b. Exposed to SUPERCLEAN for at least 6 months, started working at Glop Industries on or before September 2002
After some deliberation, you define the exposed groups as low, medium or high exposure (depending on the maximum allowable level of exposure to SUPERCLEAN for their job category) and the unexposed group as employees either not involved with SUPERCLEAN production or those working less than 6 months in SUPERCLEAN production.
You now have the basic framework of your retrospective cohort study. You have redefined your hypothesis to incorporate your assumptions about the induction period and you have clearly defined your exposure variable. You are obviously excited to get out there and begin collecting data but you must first determine who is eligible for the study.
3. How would you define eligibility criteria for study participants? [Aschengrau & Seage pp. 203-205]
4. On what would you base your definition of Susser Syndrome?[Aschengrau & Seage pp. 217-219]
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Analytic studies can be a key component of field investigations, but beware of an impulse to begin one too quickly. Studies can be time- and resource-intensive, and a hastily constructed study might not answer the correct questions. For example, in a foodborne disease outbreak investigation, if the culprit food is not on your study’s questionnaire, you probably will not be able to implicate it. Analytic studies typically should be used to test hypotheses, not generate them. However, in certain situations, collecting data quickly about patients and a comparison group can be a way to explore multiple hypotheses. In almost all situations, generating hypotheses before designing a study will help you clarify your study objectives and ask better questions.
The initial steps of an investigation, described in previous chapters, are some of your best sources of hypotheses. Key activities include the following:
In November 2014, a US surveillance system for foodborne diseases (PulseNet) detected a cluster (i.e., a possible outbreak) of listeriosis cases based on similar-appearing Listeria monocytogenes isolates by pulsed-field gel electrophoresis of the isolates. No suspected foods were identified through routine patient interviews by using a Listeria -specific questionnaire with approximately 40 common food sources of listeriosis (e.g., soft cheese and deli meat). The outbreak’s descriptive epidemiology offered no clear leads: the sex distribution was nearly even, the age spectrum was wide, and the case-fatality rate of approximately 20% was typical. Notably, however, 3 of the 35 cases occurred among previously healthy school-aged children, which is highly unusual for listeriosis. Most cases occurred during late October and early November.
Investigators began reinterviewing patients by using a hypothesis-generating shotgun questionnaire with more than 500 foods, but it did not include caramel apples. By comparing the first nine patient responses with data from a published survey of food consumption, strawberries and ice cream emerged as hypotheses. However, several interviewed patients denied having eaten these foods during the month before illness. An investigator then conducted lengthy, open-ended interviews with patients and their family members. During one interview, he asked about special foods eaten during recent holidays, and the patient’s wife replied that her husband had eaten prepackaged caramel apples around Halloween. Although produce items had been implicated in past listeriosis outbreaks, caramel apples seemed an unlikely source. However, the interviewer took note of this connection because he had previously interviewed another patient who reported having eaten caramel apples. This event underscores the importance of one person conducting multiple interviews because that person might make subtle mental connections that may be missed when reviewing other interviewers’ notes. In fact, several other investigators listening to the interview noted this exposure—among hundreds of others—but thought little of it.
In this investigation, the finding of high strawberry and ice cream consumption among patients, coupled with the timing of the outbreak during a holiday period, helped make a sweet food (i.e., caramel apples) seem more plausible as the possible source.
To explore the caramel apple hypothesis, investigators asked five other patients about this exposure, and four reported having eaten them. On the basis of these initial results, investigators designed and administered a targeted questionnaire to patients involved in the outbreak, as well as to patients infected with unrelated strains of L. monocytogenes (i.e., a case–case study). This study, combined with testing of apples and the apple packing facility, confirmed that caramel apples were the source (2). Had a single interviewer performed multiple open-ended interviews to generate hypotheses before the shotgun questionnaire, the outbreak might have been solved sooner.
As evident in public health and clinical guidelines, randomized controlled trials (e.g., trials of drugs, vaccines, and community-level interventions) are the reference standard for epidemiology, providing the highest level of evidence. However, such studies are not possible in certain situations, including outbreak investigations. Instead, investigators must rely on observational studies, which can provide sufficient evidence for public health action. In observational studies, the epidemiologist documents rather than determines the exposures, quantifying the statistical association between exposure and disease. Here again, the key when designing such studies is to obtain a relevant comparison group for the patients ( Box 7.2 ).
Because field analytic studies are used to quantify the association between exposure and disease, defining what is meant by exposure and disease is essential. Exposure is used broadly, meaning demographic characteristics, genetic or immunologic makeup, behaviors, environmental exposures, and other factors that might influence a person’s risk for disease. Because precise information can help accurately estimate an exposure’s effect on disease, exposure measures should be as objective and standard as possible. Developing a measure of exposure can be conceptually straightforward for an exposure that is a relatively discrete event or characteristic—for example, whether a person received a spinal injection with steroid medication compounded at a specific pharmacy or whether a person received a typhoid vaccination during the year before international travel. Although these exposures might be straightforward in theory, they can be subject to interpretation in practice. Should a patient injected with a medication from an unknown pharmacy be considered exposed? Whatever decision is made should be documented and applied consistently.
Additionally, exposures often are subject to the whims of memory. Memory aids (e.g., restaurant menus, vaccination cards, credit card receipts, and shopper cards) can be helpful. More than just a binary yes or no, the dose of an exposure can also be enlightening. For example, in an outbreak of fungal bloodstream infections linked to contaminated intravenous saline flushes administered at an oncology clinic, affected patients had received a greater number of flushes than unaffected patients ( 4 ). Similarly, in an outbreak of Listeria monocytogenes infections, the association with deli meat became apparent only when the exposure evaluated was consumption of deli meat more than twice a week ( 5 ).
Defining disease (e.g., does a person have botulism?) might sound simple, but often it is not; read more about making and applying disease case definitions in Chapter 3 .
Three types of observational studies are commonly used in the field. All are best performed by using a standard questionnaire specific for that investigation, developed on the basis of hypothesis-generating interviews.
Observational Study Type 1: Cohort
In concept, a cohort study, like an experimental study, begins with a group of persons without the disease under study, but with different exposure experiences, and follows them over time to find out whether they experience the disease or health condition of interest. However, in a cohort study, each person’s exposure is merely recorded rather than assigned randomly by the investigator. Then the occurrence of disease among persons with different exposures is compared to assess whether the exposures are associated with increased risk for disease. Cohort studies can be prospective or retrospective.
Prospective Cohort Studies
A prospective cohort study enrolls participants before they experience the disease or condition of interest. The enrollees are then followed over time for occurrence of the disease or condition. The unexposed or lowest exposure group serves as the comparison group, providing an estimate of the baseline or expected amount of disease. An example of a prospective cohort study is the Framingham Heart Study. By assessing the exposures of an original cohort of more than 5,000 adults without cardiovascular disease (CVD), beginning in 1948 and following them over time, the study was the first to identify common CVD risk factors ( 6 ). Each case of CVD identified after enrollment was counted as an incident case. Incidence was then quantified as the number of cases divided by the sum of time that each person was followed (incidence rate) or as the number of cases divided by the number of participants being followed (attack rate or risk or i ncidence proportion). In field epidemiology, prospective cohort studies also often involve a group of persons who have had a known exposure (e.g., survived the World Trade Center attack on September 11, 2001 [ 7 ]) and who are then followed to examine the risk for subsequent illnesses with long incubation or latency periods.
Retrospective Cohort Studies
A retrospective cohort study enrolls a defined participant group after the disease or condition of interest has occurred. In field epidemiology, these studies are more common than prospective studies. The population affected is often well-defined (e.g., banquet attendees, a particular school’s students, or workers in a certain industry). Investigators elicit exposure histories and compare disease incidence among persons with different exposures or exposure levels.
Observational Study Type 2: Case–Control
In a case–control study, the investigator must identify a comparison group of control persons who have had similar opportunities for exposure as the case-patients. Case–control studies are commonly performed in field epidemiology when a cohort study is impractical (e.g., no defined cohort or too many non-ill persons in the group to interview). Whereas a cohort study proceeds conceptually from exposure to disease or condition, a case–control study begins conceptually with the disease or condition and looks backward at exposures. Excluding controls by symptoms alone might not guarantee that they do not have mild cases of the illness under investigation. Table 7.1 presents selected key differences between a case–control and retrospective cohort study.
Feature | Retrospective cohort study | Case–control study | Case–case study |
---|---|---|---|
Sample size | Larger | Smaller | Smaller |
Costs | More (because of size) | Less | Less |
Study time | Short | Short | Short |
If disease is rare | Inefficient | Efficient | Efficient (if comparison cases already identified) |
If exposure is rare | Efficient | Inefficient | Inefficient |
If multiple exposures are relevant | Often can examine | Can examine | Can examine |
If patients have multiple outcomes | Can examine | Cannot examine | Cannot examine |
Natural history | Can ascertain | Cannot ascertain | Cannot ascertain |
Disease risk | Can measure | Cannot measure | Cannot measure |
Recall bias | Potential problem | Potential problem | Generally less of a problem |
Selection bias | Potential problem | Potential problem | Potential problem |
If population is not well-defined | Difficult | Advantageous | Advantageous |
Observational Study Type 3: Case–Case
In case–case studies, a group of patients with the same or similar disease serve as a comparison group (8). This method might require molecular subtyping of the suspected pathogen to distinguish outbreak-associated cases from other cases and is especially useful when relevant controls are difficult to identify. For example, controls for an investigation of Listeria illnesses typically are patients with immunocompromising conditions (e.g., cancer or corticosteroid use) who might be difficult to identify among the general population. Patients with Listeria isolates of a different subtype than the outbreak strain can serve as comparisons to help reduce bias when comparing food exposures. However, patients with similar illnesses can have similar exposures, which can introduce a bias, making identifying the source more difficult. Moreover, other considerations should influence the choice of a comparison group. If most outbreak-associated case-patients are from a single neighborhood or are of a certain race/ethnicity, other patients with listeriosis from across the country will serve as an inadequate comparison group.
Considerations for Selecting Controls
Selecting relevant controls is one of the most important considerations when designing a case–control study. Several key considerations are presented here; consult other resources for in-depth discussion ( 9,10 ). Ideally, controls should
Controls should be chosen independently of any specific exposure under evaluation. If you select controls on the basis of lack of exposure, you are likely to find an association between illness and that exposure regardless of whether one exists. Also important is selecting controls from a source population in a way that minimizes confounding (see Chapter 8 ), which is the existence of a factor (e.g., annual income) that, by being associated with both exposure and disease, can affect the associations you are trying to examine.
When trying to enroll controls who reflect the source population, try to avoid overmatching (i.e., enrolling controls who are too similar to case-patients, resulting in fewer differences among case-patients and controls than ought to exist and decreased ability to identify exposure–disease associations). When conducting case–control studies in hospitals and other healthcare settings, ensure that controls do not have other diseases linked to the exposure under study.
Commonly Used Control Selection Methods
When an outbreak does not affect a defined population (e.g., potluck dinner attendees) but rather the community at large, a range of options can be used to determine how to select controls from a large group of persons.
When considering outside data sources, investigators must determine whether those data provide an appropriate comparison group. For example, persons in surveys might differ from case-patients in ways that are impossible to determine. Other patients might be so similar to case-patients that risky exposures are unidentifiable, or they might be so different that exposures identified as risks are not true risks.
To help control for confounding, controls can be matched to case-patients on characteristics specified by investigators, including age group, sex, race/ethnicity, and neighborhood. Such matching does not itself reduce confounding, but it enables greater efficiency when matched analyses are performed that do ( 15 ). When deciding to match, however, be judicious. Matching on too many characteristics can make controls difficult to find (making a tough process even harder). Imagine calling hundreds of random telephone numbers trying to find a man of a particular ethnicity aged 50–54 years who is then willing to answer your questions. Also, remember not to match on the exposure of interest or on any other characteristic you wish to examine. Matched case–control study data typically necessitate a matched analysis (e.g., conditional logistic regression) ( 15 ).
Matching Types
The two main types of matching are pair matching and frequency matching.
Pair Matching
In pair matching, each control is matched to a specific case-patient. This method can be helpful logistically because it allows matching by friends or relatives, neighborhood, or telephone exchange, but finding controls who meet specific criteria can be burdensome.
Frequency Matching
In frequency matching, also called category matching , controls are matched to case-patients in proportion to the distribution of a characteristic among case-patients. For example, if 20% of case-patients are children aged 5–18 years, 50% are adults aged 19–49 years, and 30% are adults 50 years or older, controls should be enrolled in similar proportions. This method works best when most case-patients have been identified before control selection begins. It is more efficient than pair matching because a person identified as a possible control who might not meet the criteria for matching a particular case-patient might meet criteria for one of the case-patient groups.
Number of Controls
Most field case–control studies use control-to-case-patient ratios of 1:1, 2:1, or 3:1. Enrolling more than one control per case-patient can increase study power, which might be needed to detect a statistically significant difference in exposure between case-patients and controls, particularly when an outbreak involves a limited number of cases. The incremental gain of adding more controls beyond three or four is small because study power begins to plateau. Note that not all case-patients need to have the same number of controls. Sample size calculations can help in estimating a target number of controls to enroll, although sample sizes in certain field investigations are limited more by time and resource constraints. Still, estimating study power under a range of scenarios is wise because an analytic study might not be worth doing if you have little chance of detecting a statistically significant association. Sample size calculators for unmatched case–control studies are available at http://www.openepi.com and in the StatCalc function of Epi Info ( https://www.cdc.gov/epiinfo ).
More than One Control Group
Sometimes the choice of a control group is so vexing that investigators decide to use more than one type of control group (e.g., a hospital-based group and a community group). If the two control groups provide similar results and conclusions about risk factors for disease, the credibility of the findings is increased. In contrast, if the two control groups yield conflicting results, interpretation becomes more difficult.
Since the 1940s, field epidemiology students have studied a classic outbreak of gastrointestinal illness at a church potluck dinner in Oswego, New York ( 16 ). However, the case study presented here, used to illustrate study designs, is a different potluck dinner.
In April 2015, an astute neurologist in Lancaster, Ohio, contacted the local health department about a patient in the emergency department with a suspected case of botulism. Within 2 hours, four more patients arrived with similar symptoms, including blurred vision and shortness of breath. Health officials immediately recognized this as a botulism outbreak.
Descriptive epidemiology and questioning of the case-patients revealed that all had eaten at the same church potluck dinner and had no other common exposures, making the potluck the likely exposure site and attendees the likely source population. Thus, an analytic study would be targeted at potluck attendees, although investigators must remain alert to case-patients among nonattendees. As initial interviews were conducted, more cases of botulism were being diagnosed, quickly increasing to more than 25. The source of the outbreak needed to be identified rapidly to halt further exposure and illness.
In the actual outbreak, 29 (38%) of 77 potluck attendees had botulism. The investigators performed a cohort study, interviewing 75 of the 77 attendees about 52 foods served ( 17 ). The attack rate among persons who had eaten potato salad was significantly and substantially higher than the attack rate among those who had not, with a risk ratio of 14 (95% confidence interval 5–42). One of the potato salads served was made with incorrectly home-canned potatoes (a known source of botulinum toxin), and samples of discarded potato salad tested positive for botulinum toxin, supporting the findings of the analytic study. (Of note, persons often blame potato salad for causing illness when, in fact, it rarely is a source. This outbreak was a notable exception.)
In field epidemiology, the link between exposure and illness is often so strong that it is evident despite such inherent study limitations as small sample size and exposure misclassification. In this outbreak, a few of the patients with botulism reported not having eaten potato salad, and some of the attendees without botulism reported having eaten it. In epidemiologic studies, you rarely find 100% concordance between exposure and outcome for various reasons, including incomplete or erroneous recall because remembering everything eaten is difficult. Here, cross-contamination of potato salad with other foods might have helped explain cases among patients who had not eaten potato salad because only a small amount of botulinum toxin is needed to produce illness.
Two- by- two tables are covered in more detail in Chapter 8 .
Ill | Not Ill | |
---|---|---|
Exposed | a | b |
Unexposed | c | d |
A risk ratio cannot be calculated from a case–control study because true attack rates cannot be calculated.
III (Cases) | Not III (Controls) | |
---|---|---|
Exposed | a | b |
Unexposed | c | d |
What kind of study would you design if your hypothesis-generating interviews lead you to believe that everyone, or nearly everyone, was exposed to the same suspected infection source? How would you test hypotheses if all barbecue attendees, ill and non-ill, had eaten the chicken or if all town residents had drunk municipal tap water, and no unexposed group exists for comparison? A few factors that might be of help are the exposure timing (e.g., a particularly undercooked batch of barbeque), the exposure place (e.g., a section of the water system more contaminated than others), and the exposure dose (e.g., number of chicken pieces eaten or glasses of water drunk). Including questions about the time, place, and frequency of highly suspected exposures in a questionnaire can improve the chances of detecting a difference ( 18 ).
Cohort, case–control, and case–case studies are the types of analytic studies that field epidemiologists use most often. They are best used as mechanisms for evaluating—quantifying and testing—hypotheses identified in earlier phases of the investigation. Cohort studies, which are oriented conceptually from exposure to disease, are appropriate in settings in which an entire population is well-defined and available for enrollment (e.g., guests at a wedding reception). Cohort studies are also appropriate when well-defined groups can be enrolled by exposure status (e.g., employees working in different parts of a manufacturing plant). Case–control studies, in contrast, are useful when the population is less clearly defined. Case–control studies, oriented from disease to exposure, identify persons with disease and a comparable group of persons without disease (controls). Then the exposure experiences of the two groups are compared. Case–case studies are similar to case–control studies, except that controls have an illness not linked to the outbreak. Case–control studies are probably the type most often appropriate for field investigations. Although conceptually straightforward, the design of an effective epidemiologic study requires many careful decisions. Taking the time needed to develop good hypotheses can result in a questionnaire that is useful for identifying risk factors. The choice of an appropriate comparison group, how many controls per case-patient to enroll, whether to match, and how best to avoid potential biases are all crucial decisions for a successful study.
This chapter relies heavily on the work of Richard C. Dicker, who authored this chapter in the previous edition.
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Methodology
Published on February 10, 2023 by Tegan George . Revised on June 22, 2023.
A retrospective cohort study is a type of observational study that focuses on individuals who have an exposure to a disease or risk factor in common. Retrospective cohort studies analyze the health outcomes over a period of time to form connections and assess the risk of a given outcome associated with a given exposure.
It is crucial to note that in order to be considered a retrospective cohort study, your participants must already possess the disease or health outcome being studied.
When to use a retrospective cohort study, examples of retrospective cohort studies, advantages and disadvantages of retrospective cohort studies, other interesting articles, frequently asked questions.
Retrospective cohort studies are a type of observational study . They are often used in fields related to medicine to study the effect of exposures on health outcomes. While most observational studies are qualitative in nature, retrospective cohort studies are often quantitative , as they use preexisting secondary research data. They can be used to conduct both exploratory research and explanatory research .
Retrospective cohort studies are often used as an intermediate step between a weaker preliminary study and a prospective cohort study , as the results gleaned from a retrospective cohort study strengthen assumptions behind a future prospective cohort study.
A retrospective cohort study could be a good fit for your research if:
Retrospective cohort studies use secondary research data, such as existing medical records or databases, to identify a group of people with an exposure or risk factor in common. They then look back in time to observe how the health outcomes developed. Case-control studies rely on primary research , comparing a group of participants with a condition of interest to a group lacking that condition in real time.
Retrospective cohort studies are common in fields like medicine, epidemiology, and healthcare.
You collect data from participants’ exposure to organophosphates, focusing on variables like the timing and duration of exposure, and analyze the health effects of the exposure. Example: Healthcare retrospective cohort study You are examining the relationship between tanning bed use and the incidence of skin cancer diagnoses.
Retrospective cohort studies can be a good fit for many research projects, but they have their share of advantages and disadvantages.
If you want to know more about statistics , methodology , or research bias , make sure to check out some of our other articles with explanations and examples.
Research bias
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The primary difference between a retrospective cohort study and a prospective cohort study is the timing of the data collection and the direction of the study.
A retrospective cohort study looks back in time. It uses preexisting secondary research data to examine the relationship between an exposure and an outcome. Data is collected after the outcome you’re studying has already occurred.
Alternatively, a prospective cohort study follows a group of individuals over time. It collects data on both the exposure and the outcome of interest as they are occurring. Data is collected before the outcome of interest has occurred.
Retrospective cohort studies are at high risk for research biases like recall bias . Whenever individuals are asked to recall past events or exposures, recall bias can occur. This is because individuals with a certain disease or health outcome of interest are more likely to remember and/or report past exposures differently to individuals without that outcome. This can result in an overestimation or underestimation of the true relationship between variables and affect your research.
No, retrospective cohort studies cannot establish causality on their own.
Like other types of observational studies , retrospective cohort studies can suggest associations between an exposure and a health outcome. They cannot prove without a doubt, however, that the exposure studied causes the health outcome.
In particular, retrospective cohort studies suffer from challenges arising from the timing of data collection , research biases like recall bias , and how variables are selected. These lead to low internal validity and the inability to determine causality.
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Hypothesis testing in statistics helps us use data to make informed decisions. It starts with an assumption or guess about a group or population—something we believe might be true. We then collect sample data to check if there is enough evidence to support or reject that guess. This method is useful in many fields, like science, business, and healthcare, where decisions need to be based on facts.
Learning how to do hypothesis testing in statistics step-by-step can help you better understand data and make smarter choices, even when things are uncertain. This guide will take you through each step, from creating your hypothesis to making sense of the results, so you can see how it works in practical situations.
Table of Contents
Hypothesis testing is a method for determining whether data supports a certain idea or assumption about a larger group. It starts by making a guess, like an average or a proportion, and then uses a small sample of data to see if that guess seems true or not.
For example, if a company wants to know if its new product is more popular than its old one, it can use hypothesis testing. They start with a statement like “The new product is not more popular than the old one” (this is the null hypothesis) and compare it with “The new product is more popular” (this is the alternative hypothesis). Then, they look at customer feedback to see if there’s enough evidence to reject the first statement and support the second one.
Simply put, hypothesis testing is a way to use data to help make decisions and understand what the data is really telling us, even when we don’t have all the answers.
Hypothesis testing is important because it helps us make smart choices and understand data better. Here’s why it’s useful:
Here’s a simple guide to understanding hypothesis testing, with an example:
Explanation: Start by defining two statements:
Example: Suppose a company says their new batteries last an average of 500 hours. To check this:
Explanation: Pick a statistical test that fits your data and your hypotheses. Different tests are used for various kinds of data.
Example: Since you’re comparing the average battery life, you use a one-sample t-test .
Explanation: Decide how much risk you’re willing to take if you make a wrong decision. This is called the significance level, often set at 0.05 or 5%.
Example: You choose a significance level of 0.05, meaning you’re okay with a 5% chance of being wrong.
Explanation: Collect your data and perform the test. Calculate the test statistic to see how far your sample result is from what you assumed.
Example: You test 30 batteries and find they last an average of 485 hours. You then calculate how this average compares to the claimed 500 hours using the t-test.
Explanation: The p-value tells you the probability of getting a result as extreme as yours if the null hypothesis is true.
Example: You find a p-value of 0.0001. This means there’s a very small chance (0.01%) of getting an average battery life of 485 hours or less if the true average is 500 hours.
Explanation: Compare the p-value to your significance level. If the p-value is smaller, you reject the null hypothesis. If it’s larger, you do not reject it.
Example: Since 0.0001 is much less than 0.05, you reject the null hypothesis. This means the data suggests the average battery life is different from 500 hours.
Explanation: Summarize what the results mean. State whether you rejected the null hypothesis and what that implies.
Example: You conclude that the average battery life is likely different from 500 hours. This suggests the company’s claim might not be accurate.
Hypothesis testing is a way to use data to check if your guesses or assumptions are likely true. By following these steps—setting up your hypotheses, choosing the right test, deciding on a significance level, analyzing your data, finding the p-value, making a decision, and reporting results—you can determine if your data supports or challenges your initial idea.
Hypothesis testing is a way to use data to make decisions. Here’s a straightforward guide:
Hypothesis testing helps you make decisions based on data. It involves setting up your initial idea, picking a significance level, doing the test, and looking at the results. By following these steps, you can make sure your conclusions are based on solid information, not just guesses.
This approach lets you see if the evidence supports or contradicts your initial idea, helping you make better decisions. But remember that hypothesis testing isn’t perfect. Things like sample size and assumptions can affect the results, so it’s important to be aware of these limitations.
In simple terms, using a step-by-step guide for hypothesis testing is a great way to better understand your data. Follow the steps carefully and keep in mind the method’s limits.
A one-tailed test assesses the probability of the observed data in one direction (either greater than or less than a certain value). In contrast, a two-tailed test looks at both directions (greater than and less than) to detect any significant deviation from the null hypothesis.
The choice of test depends on the type of data you have and the hypotheses you are testing. Common tests include t-tests, chi-square tests, and ANOVA. You get more details about ANOVA, you may read Complete Details on What is ANOVA in Statistics ? It’s important to match the test to the data characteristics and the research question.
Sample size affects the reliability of hypothesis testing. Larger samples provide more reliable estimates and can detect smaller effects, while smaller samples may lead to less accurate results and reduced power.
Hypothesis testing cannot prove that a hypothesis is true. It can only provide evidence to support or reject the null hypothesis. A result can indicate whether the data is consistent with the null hypothesis or not, but it does not prove the alternative hypothesis with certainty.
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GLP-1 indicates glucagon-like peptide-1; SGLT2, sodium-glucose cotransporter-2.
Standardized mortality ratio weighting using a propensity score.
eTable 1. ATC codes and estimated days of exposure per unit of GLP-1 receptor agonists and SGLT2 inhibitors
eTable 2. ICD-10 and procedure codes for exclusion criteria.
eTable 3. Outcome definitions
eTable 4. Definition of psychiatric disorders for subgroup analysis of the primary outcome analysis and exclusion criteria for secondary outcome analysis of incident depression and anxiety-related disorder
eTable 5. Propensity score variables and definitions
eTable 6. Population characteristics for secondary outcome analyses of suicide death and non-fatal self-harm, and self-harm
eTable 7. Population characteristics for secondary outcome analysis of incident depression and anxiety-related disorders
eTable 8. Main, additional and sensitivity analyses of the primary outcome of suicide death in Sweden and Denmark
eTable 9. Secondary outcome analyses in Sweden and Denmark
eFigure 1. Propensity score distribution in the primary outcome analysis in Sweden
eFigure 2. Propensity score distribution in the primary outcome analysis in Denmark
eFigure 3. Weighted cumulative incidence for the composite of suicide death and non-fatal self-harm in Sweden and Denmark
eFigure 4. Weighted cumulative incidence for self-harm in Sweden and Denmark
eFigure 5. Weighted cumulative incidence for the composite of incident depression and anxiety-related disorders in Sweden and Denmark
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Ueda P , Söderling J , Wintzell V, et al. GLP-1 Receptor Agonist Use and Risk of Suicide Death. JAMA Intern Med. Published online September 03, 2024. doi:10.1001/jamainternmed.2024.4369
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Question What is the association between use of glucagon-like peptide-1 (GLP-1) receptor agonists and risk of suicide death in patients treated in routine clinical practice?
Findings In this cohort study of 298 553 adults initiating a GLP-1 receptor agonist or a sodium-glucose cotransporter-2 inhibitor nationwide in Sweden and Denmark, the incidence rate of suicide death was low, and those initiating GLP-1 receptor agonist use did not experience increased risk.
Meaning This study provides reassuring data showing that those initiating GLP-1 receptor agonist use were not at increased risk of suicide death, although the study could not assess small risk increases.
Importance Concerns have been raised regarding a link between use of glucagon-like peptide-1 (GLP-1) receptor agonists and increased risk of suicidality and self-harm.
Objective To assess the association between use of GLP-1 receptor agonists and the risk of suicide death in routine clinical practice.
Design, Setting, and Participants This active-comparator new-user cohort study used nationwide register data from Sweden and Denmark from 2013 to 2021. Adults 18 to 84 years old who initiated treatment with GLP-1 receptor agonists or the comparator sodium-glucose cotransporter-2 (SGLT2) inhibitors were included. Data were analyzed from March to June 2024.
Exposure Initiation of treatment with a GLP-1 receptor agonist or SGLT2 inhibitor.
Main Outcomes and Measures The primary outcome was suicide death recorded in the cause of death registers. Secondary outcomes were the composite of suicide death and nonfatal self-harm and the composite of incident depression and anxiety-related disorders. Using propensity score weighting, hazard ratios (HRs) with 95% CIs were calculated separately in the 2 countries and pooled in a meta-analysis.
Results In total, 124 517 adults initiated a GLP-1 receptor agonist and 174 036 initiated an SGLT2 inhibitor; among GLP-1 receptor agonist users, the mean (SD) age was 60 (13) years, and 45% were women. During a mean (SD) follow-up of 2.5 (1.7) years, 77 suicide deaths occurred among users of GLP-1 receptor agonists and 71 suicide deaths occurred among users of SGLT2 inhibitors: weighted incidences were 0.23 vs 0.18 events per 1000 person-years (HR, 1.25; 95% CI, 0.83-1.88), with an absolute difference of 0.05 (95% CI, −0.03 to 0.16) events per 1000 person-years. The HR was 0.83 (95% CI, 0.70-0.97) for suicide death and nonfatal self-harm, and the HR was 1.01 (95% CI, 0.97-1.06) for incident depression and anxiety-related disorders.
Conclusions and Relevance This cohort study, including mostly patients with type 2 diabetes, does not show an association between use of GLP-1 receptor agonists and an increased risk of suicide death, self-harm, or incident depression and anxiety-related disorders. Suicide death among GLP-1 receptor agonist users was rare, and the upper limit of the confidence interval was compatible with an absolute risk increase of no more than 0.16 events per 1000 person-years.
Glucagon-like peptide-1 (GLP-1) receptor agonists are increasingly used in the treatment of type 2 diabetes and obesity. Concerns have been raised regarding suicidality and self-harm linked to use of GLP-1 receptor agonists. 1 In July 2023, the European Medicines Agency launched an investigation into the safety signal following around 150 spontaneous reports received by the agency regarding suicidal thoughts and thoughts of self-harm potentially associated with the drug. An effect of GLP-1 receptor agonists on suicidality is plausible, as GLP-1 receptors are present in the central nervous system and GLP-1 receptor agonists have been shown to cross the blood-brain barrier. 2 , 3 Previous studies have linked bariatric surgery and weight-reduction drugs to a potentially increased risk of suicide and self-harm. 4 , 5 Conversely, it has also been suggested that GLP-1 receptor agonists may protect against depression. This hypothesis is based on studies indicating that depression and type 2 diabetes may have partly overlapping causes, including neuroinflammation, and that GLP-1 receptor agonists show neuroprotective properties. 6
We conducted a cohort study using nationwide data from Sweden and Denmark to examine the association of GLP-1 receptor agonist use with suicide death. In secondary analyses, we assessed the association of GLP-1 receptor agonist use with the composite of suicide and nonfatal self-harm, as well as the composite of incident depression and anxiety-related disorders.
We used health and administrative registers in Sweden and Denmark, including the population registers, 7 , 8 prescription drug registers, 9 , 10 national patient registers, 11 , 12 and cause of death registers, 13 , 14 all with nationwide coverage in each country. The registers provide data on demographic variables and vital status, all filled prescriptions from all pharmacies in each country, diagnoses and procedures registered during all outpatient specialist care visits and hospitalizations, and causes of death. Sweden and Denmark have similar universal health care systems and register infrastructure. 15 Using the same study protocol, we conducted the analyses in Sweden and Denmark separately and performed a meta-analysis of the 2 country-specific estimates.
The study was approved by the Swedish Ethical Review Authority. Informed consent was not needed. Ethical approval is not required for register-based research in Denmark. The study is reported according to the Strengthening the Reporting of Observational Studies in Epidemiology ( STROBE ) reporting guidelines for cohort studies.
Using an active-comparator new-user design, 16 we included all new users of either GLP-1 receptor agonists or sodium-glucose cotransporter-2 (SGLT2) inhibitors (comparator) who were 18 to 84 years old during 2013 to 2021 (eTable 1 in Supplement 1 ). Patients entered the cohort at the time of their first filled prescription for either a GLP-1 receptor agonist or an SGLT2 inhibitor during the study period. SGLT2 inhibitors were used as the comparator because this drug has no known association with suicide death and was used in similar clinical situations (predominantly as second-line or third-line glucose-lowering drugs for treatment of type 2 diabetes) during the study period, with both GLP-1 receptor agonists and SGLT2 inhibitors being recommended for patients with type 2 diabetes and high cardiovascular risk. Both drugs were primarily used for type 2 diabetes during the study period.
New use was defined as initiation of either drug among patients with no previous use of either drug at any time prior to cohort entry. Exclusion criteria were end-stage illness (severe malnutrition, cachexia, dementia, coma), history of dialysis, kidney transplant, major pancreatic disease, use of preparation of liraglutide with obesity indication before cohort entry (the comparator, SGLT2 inhibitors, are not used for this indication; liraglutide with obesity indication was defined using specific product codes), and no health care contact in the past year (to ensure a minimum level of health care contact and registration of health data), defined as neither outpatient care contact, hospital admission, nor use of any prescription drug (eTable 2 in Supplement 1 ).
The primary outcome was suicide death (primary or contributing cause of death), including both confirmed suicides and self-inflicted deaths of undetermined intent 17 , 18 (eTable 3 in Supplement 1 ). Secondary outcomes were a composite of suicide death and nonfatal self-harm recorded as a physician-assigned diagnosis during hospitalizations or outpatient hospital visits. 19 Self-harm was also analyzed as a separate outcome. Although nonfatal self-harm is known to be underreported in Denmark, 19 we used this outcome also in Denmark because underreporting was unlikely to differ between the study drugs, thus allowing for calculation of relative risk estimates. For the analyses of outcomes including self-harm, we excluded patients with self-harm within 3 months prior to cohort entry (eTable 2 in Supplement 1 ). This exclusion was done to avoid outcome misclassification: potential follow-up health care visits after a self-harm event may be registered with self-harm as the diagnosis and could be erroneously regarded as a new event in the analysis. Another secondary outcome was a composite of incident depression and anxiety-related disorders, recorded as diagnoses during hospitalizations or outpatient visits, or filled prescriptions for antidepressants (eTable 3 in Supplement 1 ). For this analysis, we further excluded patients with previous psychiatric disorders, defined as any psychiatric disease diagnosis or use of psychiatric medications at any time prior to cohort entry (eTable 4 in Supplement 1 ).
Patients were considered as exposed to the drug that they initiated at cohort entry until end of follow-up and were followed to outcome event, emigration, 5 years of follow-up, or end of study period (December 31, 2021). In each country separately, we estimated a propensity score for the probability of GLP-1 receptor agonist treatment conditional on variables at cohort entry, covering sociodemographic characteristics, calendar year of cohort entry, and medical history, including psychiatric diagnoses, recent health care contacts for psychiatric conditions, and prescription drug use (eTable 5 in Supplement 1 ). We controlled for confounding using the propensity score and standardized mortality ratio weighting to estimate the average treatment effect among the treated, as this can directly inform clinical decision-making. 20 After exclusion of patients outside of the common range of the propensity score for the 2 groups, and those at the 1% tails of the distribution of the common range (trimming), 21 , 22 the propensity score was reestimated, and those outside of the common range of the new propensity score were excluded. Reestimation of the propensity score after trimming is important because the model derived from the untrimmed population is mis-specified in the population that remains after trimming. 21 Standardized differences below 10% after weighting were considered as good balance between exposure groups. We estimated hazard ratios (HRs) using Cox regression, with days since cohort entry as the time scale. We performed a meta-analysis of the country-specific estimates, with a fixed-effect model using the method of Mantel and Haenszel. The absolute risk difference for the primary outcome was calculated as HR − 1 multiplied by the rate in the comparator group.
We performed a subgroup analysis for the primary outcome by history of psychiatric disorders (eTable 4 in Supplement 1 ), as the investigated safety signal is of most importance for treatment decisions among those at high risk. 23 We also performed subgroup analyses for those who initiated liraglutide and semaglutide, respectively. Propensity scores were reestimated for each secondary outcome and subgroup analysis.
We performed additional analyses for the primary and secondary outcomes. We applied an as-treated exposure definition, in which patients were censored at switch-to or add-on treatment with the other study drug or treatment discontinuation. Treatment duration was estimated based on the number of days covered by the filled prescriptions plus a 90-day period between prescriptions and after the last prescription. Furthermore, we restricted the follow-up to 1 year to assess whether a possible risk increase emerges shortly after treatment initiation.
Due to remaining imbalance in 1 baseline variable after weighting in Denmark (no use of other glucose-lowering drugs in the past 6 months), we adjusted for this variable in Denmark in post hoc sensitivity analysis of the primary and secondary outcomes. As the secondary outcome of incident depression and anxiety-related disorders included use of antidepressants, which can be prescribed also for other conditions, we performed a sensitivity analysis defining the outcome using only diagnoses registered during health care visits (eTable 3 in Supplement 1 ).
As the primary outcome analysis showed no statistically significant association, we calculated the E-value representing the minimum strength of association that an unmeasured confounder would need to have with both GLP-1 receptor agonist treatment and suicide death (conditional on the covariates included in the propensity score) to shift the confidence interval toward an increased risk such that it excludes the null. 24 Confidence intervals not including 1.0 were considered as showing a statistically significant risk difference. Data were analyzed from March to June 2024, and analyses were conducted using SAS, version 9.4 (SAS Institute).
In total, 124 517 users of GLP-1 receptor agonists and 174 036 users of SGLT2 inhibitors were included ( Figure 1 ). Among GLP-1 receptor agonist users, the mean (SD) age was 60 (13) years and 45% were women. The most commonly used GLP-1 receptor agonists were liraglutide (50%) and semaglutide (41%). Patient characteristics before and after weighting for Sweden and Denmark separately are summarized in Table 1 . The propensity score distributions are shown in eFigures 1 and 2 in Supplement 1 . Patient characteristics were well balanced after weighting, except for the proportion of patients in Denmark with no use of other glucose-lowering drugs in the past 6 months ( Table 1 ). The median (IQR) follow-up time in the primary outcome analysis was 2.8 (1.2-4.8) years for GLP-1 receptor agonist users and 2.1 (0.8-3.6) years for SGLT2 inhibitor users in Sweden, and 2.1 (0.8-4.7) years for GLP-1 receptor agonist users and 2.2 (0.8-3.9) years for SGLT2 inhibitor users in Denmark. Combining the 2 countries, mean (SD) follow-up was 2.5 (1.7) years: 2.7 (2.3) years for GLP-1 receptor agonists and 2.3 (1.6) years for SGLT2 inhibitors.
During follow-up, 77 GLP-1 receptor agonist users and 71 SGLT2 inhibitor users died by suicide. The weighted incidence rate was 0.23 vs 0.18 events per 1000 person-years (HR, 1.25; 95% CI, 0.83-1.88), with an absolute difference of 0.05 (95% CI, −0.03 to 0.16) events per 1000 person-years ( Table 2 and Figure 2 ). The country-specific HR was 1.44 (95% CI, 0.87-2.37) for Sweden and 0.94 (95% CI, 0.46-1.91) for Denmark. In the subgroup analysis, the HR was 1.25 (95% CI, 0.77-2.02) for those with a history of psychiatric disorders and 1.44 (95% CI, 0.71-2.92) for those without such a history. The HR was 1.35 (95% CI, 0.85-2.15) for those initiating liraglutide and 0.74 (95% CI, 0.33-1.67) for those initiating semaglutide ( Table 2 ).
After exclusion of those with nonfatal self-harm within 3 months prior to cohort entry, 124 459 users of GLP-1 receptor agonists and 173 985 users of SGLT2 inhibitors were included in the analysis of self-harm (patient characteristics are summarized in eTable 6 in Supplement 1 ). The HR was 0.83 (95% CI, 0.70-0.97) for the composite outcome of suicide death and nonfatal self-harm and 0.77 (95% CI, 0.65-0.91) for self-harm ( Table 3 and eFigures 3 and 4 in Supplement 1 ).
After exclusion of those with previous psychiatric disorders, 72 420 users of GLP-1 receptor agonists and 111 083 users of SGLT2 inhibitors were included in the analysis of incident depression and anxiety-related disorders (patient characteristics are summarized in eTable 7 in Supplement 1 ). The HR was 1.01 (95% CI, 0.97-1.06) ( Table 3 and eFigure 5 in Supplement 1 ). Results for Sweden and Denmark separately are summarized in eTable 8 (primary outcome) and eTable 9 (secondary outcomes) in Supplement 1 .
In the analyses of suicide death, the HR was 1.27 (95% CI, 0.76-2.15) when using the as-treated exposure definition (mean [SD] follow-up time was 1.7 [1.5] years for GLP-1 receptor agonists and 1.5 [1.3] years for SGLT2 inhibitors) and 1.11 (95% CI, 0.54-2.28) when restricting the follow-up to the first year ( Table 2 ). The post hoc sensitivity analysis, in which the variable no use of other glucose-lowering drug in the past 6 months was adjusted for, yielded an HR consistent with that of the main analysis in Denmark (eTable 8 in Supplement 1 ). The additional analyses for the secondary outcomes and the sensitivity analysis in which the secondary outcome of incident depression and anxiety-related disorders was restricted to diagnoses registered during health care visits are summarized in Table 3 . The E-value for the primary analysis showed that an unmeasured confounder would need to be associated with both GLP-1 receptor agonist use and suicide death by a risk ratio of at least 1.7 to shift the confidence interval to exclude the null.
In this cohort study of nationwide data from 2 countries, we found no statistically significant increased risk of suicide death for GLP-1 receptor agonists vs SGLT2 inhibitors used predominantly for type 2 diabetes. The upper limit of the confidence interval was compatible with up to an 88% relative-risk increase of suicide death, which corresponded to an absolute risk increase of no more than 0.16 per 1000 person-years. The findings indicate that the absolute risks of suicide death in broad groups of patients using GLP-1 receptor agonists are low and any potential risk increase would be small. We also found a slightly lower risk of self-harm and no statistically significant association with incident depression and anxiety-related disorders.
The investigated safety concern for GLP-1 receptor agonist use was based on spontaneous reports of suicidal thoughts and thoughts of self-harm. 1 To cover a wider range of conditions and outcomes associated with thoughts of self-harm and suicide, 25 , 26 we assessed suicide death, suicide death and nonfatal self-harm, and self-harm and incident depression and anxiety-related disorders. In randomized clinical trials of GLP-1 receptor agonists, increases in suicidality, depression, anxiety, and other adverse mental health outcomes have not been observed. 27 , 28 However, many trials excluded those at highest risk of suicidality through exclusion criteria or based on investigators’ judgment such that uncertainty remained regarding the generalizability of the findings to broader patient groups, and the statistical power was low due to few events. For example, a meta-analysis of clinical trials of GLP-1 receptor agonists included only 18 events of suicidal behavior among exposed patients. 28
A previous study 29 using electronic health records from the TriNetX Analytics Network assessed diagnoses of suicidal ideation during 6 months after treatment initiation with the GLP-1 receptor agonist semaglutide. In separate analyses, the study included 52 783 patients with overweight or obesity and 27 282 patients with type 2 diabetes who were prescribed semaglutide, then propensity score matched them with users of non–GLP-1 receptor agonist antiobesity or glucose-lowering medications. Semaglutide use was associated with very large relative reductions in suicidal ideation diagnosis of around 70% in both sets of analyses, indicating that the study design likely introduced important biases. For example, immortal time bias 30 may have been introduced as a new-user design was not used: patients with semaglutide prescriptions were first included regardless of their comparator drug use history, and the comparator group was selected from the remaining patients prescribed the comparator drug (ie, the exposure status for the comparator group was conditioned on not being prescribed semaglutide after cohort entry). Moreover, in the analyses of patients with type 2 diabetes, any other glucose-lowering drugs, including those used as first-line therapies, were included as the comparator, although this may have led to time-lag bias through misalignment in disease progression between the groups. 31 Furthermore, suicide death during follow-up could not be accounted for as data on vital status and causes of death were not available.
This study expands the knowledge regarding the safety of GLP-1 receptor agonists by providing data about the risk of suicide death associated with use of the drugs using an active-comparator new-user design, 16 which avoids time-related biases 31 and aligns patients in the exposed vs control group with respect to disease stage. Moreover, confounders were adjusted for using a propensity score, including a wide range of patient characteristics, and the nationwide registers enabled analysis of national study populations with virtually complete data on cause of death. 13 , 14 This study supports the conclusions of the European Medicines Agency’s investigation and the US Food and Drug Administration’s preliminary evaluation 32 that the available evidence does not support a causal association between use of GLP-1 receptor agonists and suicidal and self-injurious thoughts and actions. 33
This study has limitations. We adjusted for potential confounders, including psychiatric disorders and socioeconomic status recorded in the registers, but unmeasured confounding may have affected the results. During the study period, GLP-1 receptor agonists were predominantly used for treatment of type 2 diabetes. The drugs are increasingly used also in patients with obesity who do not have diabetes, and the study findings might not be generalizable to this group of patients. Although it is possible that some patients included in the study used GLP-1 receptor agonists off label for weight reduction, this proportion is likely to be low; a Danish study showed that the proportion of new users of semaglutide with no evidence of a type 2 diabetes diagnosis in the register data (a glycated hemoglobin value in the diabetic range, previous use of glucose-lowering medications, or diagnosis during hospital contacts) was 1% to 5% in 2018 to 2020 and 15% in 2021. 34 Moreover, in the present analyses, liraglutide (50%) and semaglutide (41%) were the most used GLP-1 receptor agonists; the subgroup analyses for each of the 2 GLP-1 receptor agonists had low numbers of events, and associations with suicidality could differ between individual drugs. Furthermore, mean follow-up time for GLP-1 receptor agonist users was 2.7 years, and although 25% of the GLP-1 receptor agonist users were followed for at least 4.7 years, risks may emerge with longer-term use. Although subgroup analyses with low numbers of events should be interpreted with much caution, especially when the subgroup hypothesis was not formulated a priori based on a suggested mechanism, 35 the HR for suicide death in Sweden was 1.44 (95% CI, 0.87-2.37), with this nominal risk difference emerging around 3 years after cohort entry.
Some suicide deaths could also be misclassified. 36 In an assessment of the reliability of the cause of death register from 2008, the proportion of suicide deaths with intent confirmed by experts was 81% in Sweden and 90% in Denmark, with few accidents and natural deaths being reclassified as suicides. 36 We also included self-inflicted deaths of undetermined intent in the primary outcome definition, as an investigation by the National Center for Suicide Research and Prevention in Sweden showed that 20% of suicide cases in the country were coded as of undetermined intent and 70% to 75% of self-inflicted deaths with unknown intent were reclassified as suicide after further investigation. 18 The overall rates of suicide death in the study population were largely in line with national estimates in both Denmark 37 and Sweden. 38 Nonfatal self-harm is known to be underreported in Denmark, 19 and the absolute risks of the secondary outcome analysis, including this outcome, are likely underestimated, as is demonstrated by the substantially lower rates observed in Denmark compared to Sweden. We could also not assess suicidal thoughts and self-harm that did not result in suicide death or a registered diagnosis during contact with the health care system. The low risk of suicide death meant that the study had limited power to assess smaller risk increases.
In this binational cohort study including predominantly patients with type 2 diabetes, use of GLP-1 receptor agonists compared with SGLT2 inhibitors was not associated with an increased risk of suicide death, self-harm, or incident depression and anxiety-related disorders. While reassuring, the study could not rule out smaller absolute risk differences for suicide death, and future studies with more outcome events should be performed as data accumulate.
Accepted for Publication: July 3, 2024.
Published Online: September 3, 2024. doi:10.1001/jamainternmed.2024.4369
Open Access: This is an open access article distributed under the terms of the CC-BY License . © 2024 Ueda P et al. JAMA Internal Medicine .
Corresponding Author: Peter Ueda, MD, PhD, Division of Clinical Epidemiology, Department of Medicine, Solna, Karolinska Institutet, 17176 Stockholm, Sweden ( [email protected] ).
Author Contributions: Drs Ueda and Pasternak had full access to all of the data in the study and take responsibility for the integrity of the data and the accuracy of the data analysis.
Concept and design: Ueda, Wintzell, Pasternak.
Acquisition, analysis, or interpretation of data: All authors.
Drafting of the manuscript: Ueda.
Critical review of the manuscript for important intellectual content: All authors.
Statistical analysis: Söderling, Wintzell.
Obtained funding: Ueda, Pasternak.
Administrative, technical, or material support: Wintzell, Melbye, Pasternak.
Supervision: Pasternak.
Conflict of Interest Disclosures: Dr Ueda reported grants from the Karolinska Institutet and the Strategic Research Area Epidemiology program during the conduct of the study. Dr Svanström reported being a former employee of IQVIA. Dr Pazzagli was supported by a grant from the Swedish Research Council for Health, Working Life, and Welfare. Dr Eliasson reported support from Konung Gustaf V:s and Drottning Victorias Frimurarstiftelse and personal fees from Novo Nordisk, Sanofi, Eli Lilly, Amgen, AstraZeneca, Boehringer Ingelheim, and Abbott outside the submitted work. Dr Melbye was supported by a grant from the Danish Cancer Society. Dr Hviid reported grants from the Novo Nordisk Foundation, the Lundbeck Foundation, and the Denmark Independent Research Fund outside the submitted work. Dr Pasternak reported grants from the Karolinska Institutet during the conduct of the study. No other disclosures were reported.
Funding/Support: This study was supported by the Strategic Research Area Epidemiology program at Karolinska Institutet.
Role of the Funder/Sponsor: Karolinska Institutet had no role in the design and conduct of the study; collection, management, analysis, and interpretation of the data; preparation, review, or approval of the manuscript; and decision to submit the manuscript for publication.
Data Sharing Statement: See Supplement 2 .
Lipids in Health and Disease volume 23 , Article number: 278 ( 2024 ) Cite this article
Metrics details
The health effects of Life’s Essential 8 (LE8) on chronic diseases have been disclosed, but its association with hypertension remains unknown. The current study aimed to explore the potential link between 10-year LE8 trajectory and the incidence of hypertension.
LE8 was constructed from four behaviors and four metabolic factors, ranging from 0 to 100. Latent mixture models were used to identify trajectories of LE8 scores during 2006 to 2016. Incident hypertension was diagnosed based on self-reported clinical diagnoses and physical examinations from 2016 to 2020. Cox models were employed to assess the association of LE8 trajectories with hypertension. In addition to incorporating the mean hs-CRP levels from 2006 to 2016, age, sex, monthly income, educational level, and occupation at recruitment were adjusted for as confounding factors.
7500 participants aged 40.28 ± 10.35 years were included in the study, of whom 2907 (38.76%) were women. Five LE8 trajectory patterns were identified. After around four-year follow-up, 667 hypertension events were observed. Compared to the Low-Stable trajectory, the hazard ratios and 95% confidence intervals for the Moderate-Increasing, Moderate-Decreasing, Moderate-Stable, and High-Stable trajectories were 0.51 (0.40, 0.65), 0.81 (0.64, 1.02), 0.45 (0.36, 0.58), 0.23 (0.16, 0.33), respectively. The risk of incident hypertension decreased as participants improved their LE8 status. The robustness of the primary results was confirmed through several sensitivity analyses.
LE8 trajectories were associated with the incident hypertension. People who improved their LE8 scores over time experienced a decreased risk of hypertension, even if they started with lower LE8 scores initially.
Hypertension accounted for 10.8 million deaths worldwide and 2.6 million deaths in China, according to the Global Burden of Disease Study 2019 [ 1 ]. Blood pressure (BP) is the most common contributing factor for cardiovascular diseases (CVD) [ 2 , 3 ]. Hypertension is a leading risk factor for all-cause mortality both globally and in China [ 4 ]. Therefore, it is imperative to explore novel strategies to reduce the health burden and promote longevity.
The American Heart Association (AHA) initially proposed Life’s Simple 7 (LS7), which includes diet, body mass index (BMI), physical activity (PA), smoking, BP, blood lipids (BL), and fasting blood glucose (FBG), as a tool to assess cardiovascular health (CVH) and to encourage its measurement, thereby advancing the promotion of CVH [ 5 ]. Recognizing the importance of sleep quality as a crucial behavior for CVH and the need for more precise CVH measurement, the AHA developed Life’s Essential 8 (LE8), which incorporated sleep quality for a more comprehensive assessment of CVH. Each factor of LE8 ranges from 0 to 100 [ 6 ]. These distinctions allow the public to better understand and quantify their CVH levels using LE8 and suggest that LE8 may serve as a more reliable and precise indicator of CVH compared to LS7. Furthermore, the potential value of LE8 in evaluating hypertension risk among the normotensive population is barely explored.
Besides its association with CVD, LE8 had also been linked to various other diseases [ 7 , 8 , 9 , 10 , 11 ] and incident hypertension [ 12 ]. However, none of these studies described the 10-year trajectory of LE8 or explored the potential benefits of improving LE8 among normotensive individuals using repeated measurements. Therefore, the longitudinal association of LE8 with incident hypertension was evaluated using data from the biennially followed Kailuan cohort study.
The Kailuan study began in 2006–2007 in the Kailuan community in Tangshan, China. It was strategically designed to elucidate potential health determinants [ 13 , 14 , 15 ]. A total of 101,510 adults were recruited between June 2006 and October 2007. Each participant underwent thorough assessments, which primarily encompassed three sections: a standardized questionnaire survey, clinical examinations, and laboratory testing. These assessments were conducted biennially.
The present study estimated the effect of the 10-year trajectory of LE8 on incident hypertension. Initially, 48,521 participants were excluded: 44,655 due to existing hypertension and 3866 due to incomplete LE8 at baseline. Additionally, 45,489 participants with new-onset hypertension or those who were lost to follow-up during the trajectory construction from 2006 to 2016 were excluded, as the goal of this study was to observe hypertension cases from 2016 to 2020, and hypertension was the outcome event that should be excluded during trajectory construction according to previous studies [ 13 ]. Finally, 7500 participants who were free of hypertension after the trajectory construction period were included (Fig. 1 ).
Flow diagram for participants included in the study
Medical workers conducted face-to-face interviews to gather demographic factors (such as sex, birthdate, level of education, monthly income, and occupation), health conditions (including self-reported clinical diagnoses of chronic diseases, and medication history), and lifestyle information (such as smoking habits, PA levels, duration of sleep, daily diet, and alcohol consumption) using a standardized questionnaire. Educational levels were determined according to the number of years of schooling, with categories including primary school or below (≤ 6 years), middle school (7–9 years), and high school or above (> 9 years). Monthly income was self-reported and categorized as < 800 CNY or ≥ 800 CNY. Occupation types were classified based on responses to questions about employment in mining and engagement in manual or mental labor. Participants were categorized as coal miners if they worked in the mine and mainly performed manual labor. Those working in the mine but engaged in mental labor were classified as other blue-collar workers. Participants not working in the mine but mainly involved in manual labor were also categorized as other blue-collar workers, while those primarily engaged in mental labor and not working in the mine were classified as white-collar workers. Participants were asked about clinically diagnosed conditions, including hypertension and other chronic diseases, with follow-up questions about medication usage if conditions were affirmed. Smoking habits were assessed by inquiring whether participants currently smoked or had quit smoking, along with detailed questions about smoking frequency. PA levels were assessed based on weekly frequency and duration of sessions lasting at least 20 min. Participants reported their average nightly sleep duration in hours. Dietary habits were evaluated through questions about flavor preferences, frequency of consuming fatty foods, and tea consumption. Alcohol consumption was categorized into three groups: current, past, and never, with detailed information gathered from participants who reported current or past drinking, including drinking patterns and withdrawal behaviors. Interviewers recorded participants’ responses to each question.
The weight and height of each participant were measured by trained medical workers, and weight (kg) divided by height squared (m 2 ) was used for BMI (kg/m 2 ) calculation. The procedures for the measurement of biochemical indicators were as follows: venous blood samples were extracted from the antecubital vein by a professional nurse following an overnight fast of about eight to twelve hours. All the samples were stored at -80 °C. Subsequently, an auto-analyzer machine (Hitachi 747; Hitachi, Tokyo, Japan) was utilized to measure biochemical indices, including FBG, total cholesterol (TC), HDL-C, and high-sensitivity C-reactive protein (hs-CRP) in these blood samples at the central laboratory of Kailuan Hospital [ 14 , 16 ]. Non-HDL-C levels were calculated as TC subtracting HDL-C. BP metrics were assessed three times with a calibrated mercury sphygmomanometer while interviewees were seated after a 5-minute rest period. BP was re-assessed if the difference between adjacent two measurements was more than five mmHg. The mean of the three BP measurements was used for subsequent analysis [ 17 , 18 ].
The LE8 primarily included four health-related indicators (BMI, FBG, BL, and BP) and four lifestyle factors (tobacco smoking, diet, sleep duration, and PA). Each component of LE8 was assessed biennially. Detailed definitions and scoring methodologies for the components of LE8 were outlined in Table S1 and elsewhere [ 19 , 20 , 21 ]. The LE8 score was counted by summing the scores of the each factor and then dividing the total score by eight. This resulted in a LE8 score ranging from 0 to 100. The LE8 trajectories in the current study depicted different change models of LE8 from 2006 to 2016 utilizing a latent mixture model.
In each follow-up survey, the interviewees were asked whether they had been clinically diagnosed with hypertension. If the answer was affirmative, further inquiries to obtain detailed information on antihypertensive medication intake. The procedure for BP measurement was explained in a previous section. Hypertension was determined based on one of the following criteria: self-reported clinical diagnosis, intake of antihypertensive drugs, or increased systolic/diastolic BP (≥ 140/90 mm Hg) [ 22 ].
The potential confounders were selected based on previous studies [ 23 , 24 ]. Demographic information, including age, sex, monthly income, educational status, and occupation, was included. Alcohol consumption was categorized as never, past, or current based on information acquired during the interview. Given the significant impact of inflammation on the cardiovascular system [ 25 ], the mean hs-CRP level during the period from 2006 to 2016 was also included as an important covariate. Apart from hs-CRP, the other covariates were collected in 2006 and incorporated into subsequent analyses according to a previous study [ 13 ].
Continuous variables were described using mean ± standard deviation and median with interquartile range (25–75%), while categorical variables were presented as numbers and percentages. The person-time was calculated from the 2016 survey until the onset of hypertension, death, loss to follow-up, or the endpoint (December 31, 2020), whichever came first.
A latent mixture model within the PROC TRAJ procedure was employed to classify the trajectory patterns of LE8 from 2006 to 2016, following established methodology [ 26 , 27 ]. Various parameters and fit statistics were used to determine the optimal number of trajectories, including the proportion in each trajectory, Bayesian Information Criteria, average posterior probability in each trajectory, and visual inspection of the trajectories. The model with five trajectories provided the best fit (Table S2 ). Models with different functional forms (e.g., 3-cubic, 2-quadratic, and 1-linear) were compared and evaluated based on their significance levels, utilizing the highest polynomial term. Ultimately, we identified two pattern with quadratic order terms and three patterns with up to cubic order terms, which met the statistical conditions necessary for constructing trajectory patterns. The Low-Stable trajectory was set as the reference for subsequent analysis.
The cumulative incidence of hypertension by LE8 trajectories was described using the Kaplan-Meier method, followed by a Log-rank test. The association between these LE8 trajectory patterns and the risk of hypertension was evaluated using the Cox model to calculate hazard ratios (HR) and 95% confidence intervals (CI). Before constructing the model, the proportional hazards assumption was satisfied using Schoenfeld residual methods [ 28 ]. The crude model was adjusted for none, while Model 1 was adjusted for age (years), sex (women, men), education level (elementary school or below, middle school, high school and above), monthly income (< 800, ≥ 800 CNY), occupation (coal miner, other blue-collar, white collar), alcohol consumption (never, abstainer, current), and mean hs-CRP concentration during 2006–2016. Additionally, LE8 scores in 2006, 2008, 2010, 2012, 2014, and 2016 were included as additional adjustments based on Model 1 to examine whether the association of LE8 trajectories with incident hypertension could be explained by a single LE8 measurement in the follow-up.
The association between the slope of LE8 during 2006–2016 and hypertension risk was examined. The magnitude of the annual change rate indicates the extent of the annual change, with a larger magnitude indicating a greater rate of change and a steeper trend for LE8. The positive or negative symbol associated with the slope indicates the overall direction of change: a positive slope denotes an increase, while a negative slope represents a decrease. The slope calculation methods were in accordance with a previous study [ 29 ]. A restricted cubic spline model with three (25th, 50th, 75th percentiles) knots was performed to evaluate the underlying relationship between LE8 slope and incident hypertension after adjusting for the impact of confounders and the 2006 LE8. This approach helps capture potential non-linear associations between the annual change rate of LE8 and hypertension risk.
Note: x: Timepoint of Measurement y: Life’s Essential 8 Slope: Annual Change Rate
Several sensitivity analyses were conducted. First, the primary results were re-analyzed after excluding women, considering the distribution of sex and the particularity of occupation. Second, participants were divided into five groups according to the quintile of 2016 LE8, and numeric values were assigned to the LE8 groups to examine the trend in the adjusted model. Third, stratified and interaction analyses were performed for age (< 45, ≥ 45) and sex (women, men). Fourth, to account for differences in gender and age among the five trajectory groups, the strata option was used for sex and age modeled as a time scale, allowing for better control of the impact of these factors. Fifth, BP was removed from LE8, and then the LE8 score was recalculated using the remaining seven factors to exclude the impact derived from BP. Sixth, to address the issue of multiple comparisons and potential inflation of type-I error rates, the Bonferroni correction was applied by adjusting alpha and the corresponding CIs. The alpha level was set as 0.05/4, as there were four hypothesis tests conducted. Seventh, the ‘evalue hr’ code in Stata was used to calculate E-values to consider the potential impact derived from unmeasured residual factors [ 30 ]. The comparison between the inclusion and exclusion groups was conducted because the non-participant population was significantly larger than the participant population. Individuals who self-reported clinically diagnosed hypertension were further excluded to address potential errors in self-reported hypertension.
All analyses were conducted using Stata/MP2 V17 (Stata Corp LLC, Texas, TX, United States) and SAS 9.4 (SAS Institute, Cary, NC), and a 2-sided P value < 0.05 was deemed significant.
A total of 7500 individuals were included in the analyses, all of whom met the inclusion criteria and were free of hypertension from 2006 to 2016. The mean (SD) age at baseline in 2016 was 40.28 (10.35) years, with 2,907 (38.76%) of the participants being women. Five LE8 trajectories from 2006 to 2016 were determined: Low-Stable ( n = 762), Moderate-Increasing ( n = 1275), Moderate-Decreasing ( n = 1172), Moderate-Stable ( n = 2690), and High-Stable ( n = 1601) (Fig. 2 ). Individuals in the High-Stable trajectory tended to be women, younger, engaged in other blue-collar occupations, abstainers from alcohol, and had higher educational statuses (Table 1 ).
Mean Life’s Essential 8 score from 2006 to 2016 according to five Life’s Essential 8 trajectories. High-Stable (brown line) means participants maintained a high LE8 score. Moderate-Stable (magenta line) means participants maintained a moderate LE8 score. Moderate-Decreasing (orange line) means participants started with a moderate LE8 score and then decreased. Moderate-Increasing (blue line) means participants started with a moderate LE8 score and then increased. Low-Stable (purple line) means participants maintained a low LE8 score
The incidence of hypertension significantly varied across trajectories ( P for log-rank test < 0.001) (Figure S1 ). In the trajectory analysis, a total of 667 incident hypertension events were diagnosed during the period from 2016 to 2020. LE8 trajectories were significantly associated with new-onset hypertension (Table 2 ). Compared to the Low-Stable trajectory, the Moderate-Increasing (adjusted HR: 0.51; 95% CI, 0.40, 0.65), Moderate-Deceasing (adjusted HR: 0.81; 95% CI, 0.64, 1.02), Moderate-Stable (adjusted HR: 0.45; 95% CI, 0.36, 0.58), and High-Stable (adjusted HR: 0.23; 95% CI, 0.16, 0.33) trajectories were all related to a lower risk of hypertension. These results remained stable after additional adjustments for LE8 scores in 2006, 2008, 2010, 2012, 2014, and 2016 as continuous variables. Improvement in LE8 was related to a decreased incident hypertension risk, and the faster the growth, the more the risk falls (Fig. 3 ).
Association between annual change rate of Life’s Essential 8 and incident hypertension. The y-axis is hazard ratio with the area within dotted line representing 95% CIs. Data were fitted by a restricted cubic spline Cox proportional hazards regression model adjusted for age (years), sex (women, men), educational level (elementary school or below, middle school, high school and above), income per month (< 800, ≥ 800 CNY), occupation (coal miner, other blue-collar, white collar), alcohol consumption (never, abstainer, current), mean serum concentration of hs-CRP during 2006–2016, and 2006 Life’s Essential 8 measurement
After excluding female participants, the adjusted HR comparing the High-Stable trajectory with the Low-Stable one was 0.28 (95% CI, 0.17, 0.46) (Table S3 ). Consistently, a higher LE8 level was related to a lower risk of incident hypertension ( P -trend < 0.001) (Table S4 ). Moreover, the results from stratified analysis further supported the primary findings, and no effect-modifying factors were found in the association between LE8 trajectories and incident hypertension ( P -interaction > 0.05 for all) (Table S5 ). The results remained virtually unchanged when age was modeled as a time scale and sex was used as the strata option (Table S6 ). LE7, removing BP from LE8, was strongly associated with the risk of incident hypertension, and the results were consistent with the primary findings (Table S7 ). Slight changes were observed after Bonferroni correction (Table S8 ). The E-value tested the sensitivity to unmeasured confounding, and the primary findings remained stable, suggesting that they may not be invalidated with the current evidence (Table S9 ). Excluded participants were older, had relatively lower LE8 levels, higher systemic inflammation, were predominantly men, and had lower educational attainment compared to those included (Table S10 ). The results remained consistent with the primary findings after accounting for potential errors in self-reported hypertension (Table S11 ).
Five distinct trajectories were identified, which revealed varying associations with the risk of hypertension. Individuals who consistently maintained the highest LE8 level over the 10-year period experienced a remarkable 78% reduction in the risk of developing incident hypertension compared to those who consistently remained at the lowest LE8 level. The findings suggested that improvement in LE8 status was beneficial for hypertension prevention, regardless of baseline LE8 status, and indicate that the faster the improvement, the better.
The present study included 7500 participants who were free of hypertension from 2006 to 2010, representing a subset of the original Kailuan cohort. This selection was made due to the high incidence of hypertension and the restrictive inclusion criteria necessary for the current research objectives. The included participants were not fully representative, as they were generally healthier than those excluded. Specifically, they were younger, had a favorable sex distribution, and possessed higher levels of education. These factors naturally positioned them at an advantageous status for hypertension prevention, making them less likely to develop hypertension. Although the current study may have underestimated the actual association between LE8 trajectories and incident hypertension, a significant association between LE8 trajectories and incident hypertension persisted even after adjusting for each LE8 measurement from 2006 to 2010. Research unveiled the health effects of LE8 on various adverse health conditions [ 13 , 14 , 15 , 16 ]. However, prior studies primarily focused on individual health behaviors within LE8 and their associations with incident hypertension [ 31 ]. Another study involving 52,990 participants demonstrated that a higher LE8 score is linked to a reduced risk of hypertension [ 12 ]. Yet, the health effects of the 10-year trajectory of LE8 and its improvement on hypertension remained unexplored. Recently, a study indicated that a higher LE8 was linked to a reduced risk of hypertension. [ 32 ]. Another study delved into the association of four health behaviors within LE8 with the incidence of hypertension, revealing that each health behavior was independently associated with incident hypertension, and improvement in any of these behaviors was related to a reduced risk of developing hypertension [ 31 ]. The Jackson Heart Study also revealed that unhealthy lifestyles, characterized by less PA, smoking, and an absence of an ideal diet, were linked to a heightened risk of hypertension [ 33 ]. Additionally, sleep duration, as a new factor added to LE8, has been reported to be related to the risk of hypertension, indicating that poor sleep quality is associated with new-onset hypertension [ 34 ]. However, these studies failed to comprehensively investigate the association of complete LE8 with hypertension and discuss the long-term pattern of LE8 and its association with hypertension.
The current findings contributed to the body of evidence suggesting an inverse association between LE8 and the risk of hypertension. Importantly, despite the relatively low LE8 status at baseline, the improvement in LE8 significantly attenuated the risk of hypertension significantly (HR: 0.61, 95% CI: 0.48, 0.78 for Moderate-Increasing trajectory; HR: 0.79, 95% CI: 0.62, 0.99 for Moderate-Decreasing trajectory). These results underscored a critical concern, emphasizing the effectiveness of enhancing healthy behaviors and metabolic status within the context of LE8 as a preventive measure against the development of hypertension. The findings from a previous large-scale study revealed that arterial stiffness status partially mediated the association between LE8 and stroke, suggesting that the stiffness of blood vessels may serve as a potential mechanism [ 35 ]. These results underscored the importance of robust public health efforts aimed at improving CVH, which may counterbalance arterial stiffness. In essence, despite an overall decreasing trend in LE8 over time, individuals still have the potential to enhance their LE8 status. Regardless of whether they had lower LE8 levels at the beginning or not, they can benefit from improving their LE8 status. This highlighted the significance of ongoing efforts to promote healthy behaviors and metabolic health, as they may contribute to maintaining or improving CVH and potentially mitigate the risk of hypertension.
In contemporary times, individuals with ideal LE8 scores were relatively scarce [ 36 , 37 , 38 ]. However, the benefits of achieving a high LE8 extended beyond CVH, impacting a wide range of organs and systems [ 13 , 14 , 15 , 16 ]. The rate of improvement in LE8 over time emerges as a critical factor in reducing the risk of hypertension, with lower hypertension risk observed in individuals experiencing faster LE8 increases. These findings underscored the importance of continuous improvement in LE8 for reducing hypertension risk and enhancing CVH. LE8 may serve as an effective and feasible tool for evaluating the future risk of hypertension and identifying populations at high risk. This, in turn, could guide individuals towards addressing their health deficiencies, potentially preventing hypertension and its associated complications. Moreover, the utilization of LE8 could inform health managers in crafting timely and real-time health policies for residents, enhancing public health outcomes. Given the costs associated with LE8 measurement, alternative strategies such as implementing cost-effective campaigns such as health education and health knowledge contests could encourage the public to optimize their CVH status. Overall, striving to attain higher LE8 scores could prove to be a valuable strategy, offering a substantial reduction in hypertension risk and improving overall CVH in the population.
Some studies derived from the China Health and Retirement Longitudinal Study (CHARLS), a nationally representative survey of individuals aged 45 and older in China, reported that approximately 25% of participants developed hypertension from 2011 to 2018 [ 39 , 40 , 41 ]. Although this rate was higher than the 8.89% incidence observed in the current study, it was difficult to draw a definitive conclusion. Because the lower incidence in this study may be attributed to the fact that participants were free of hypertension for the previous ten years, which could introduce a healthy population effect and attenuate the association between LE8 and incident hypertension. However, the significantly inverse association between LE8 and the risk of hypertension in this study remained, which further emphasized the health benefits of enhancing LE8.
Research conducted within the Atherosclerosis Risk in Communities study revealed a significant association between improvement in LS7 from midlife to late life and enhanced systolic and diastolic function, along with improvements in cardiovascular structure and function [ 42 , 43 ]. Moreover, a Chinese cohort study involving more than 40,000 individuals demonstrated a potential role of LE8 in mitigating atherosclerosis progression [ 35 ]. Additionally, a study reported a noteworthy association between a higher cardiovascular score and decreased concentrations of pro-atherosclerotic and cardiac stress factors, as well as neurohormonal biomarkers, while simultaneously exhibiting higher levels of cardioprotective biomarkers [ 44 ]. Overall, these studies provided valuable insights into the biological pathways through which improvements in LE8 and CVH may influence cardiovascular structure, function, and disease progression, further highlighting the importance of optimizing CVH to reduce the risk of hypertension and improve overall cardiovascular outcomes.
LE8 data were collected repeatedly from 2006 to 2016, ensuring consistency and minimizing random error. Moreover, this study was the first to discuss the impact of 10-year trajectories of LE8 and incident hypertension. However, several limitations should be acknowledged. First, the Kailuan cohort was predominantly composed of men, primarily representing people from the Kailuan community, potentially limiting the generalizability of current findings. Second, while efforts were made to account for potential confounders, some variables remain unmeasured. This limitation was mitigated by calculating an E-value to assess their potential impact and verify the stability of current findings. Third, the 10-year duration of LE8 trajectories examined in the current study may not entirely reflect an individual’s CVH throughout their lifespan. Fourth, the current results relied heavily on questionnaires, which could introduce biases and reduce reliability. To address this issue, future studies should consider quantifying LE8 components where feasible. For example, biomarkers such as urinary cotinine could be utilized to quantify smoking status, thereby enhancing the accuracy of LE8 assessments. Fifth, the complex interplay of factors such as alcohol consumption with obesity, diabetes, and smoking suggests careful consideration. Despite removing BP from LE8 calculation and conducting sensitivity analysis to examine the stability of the current findings, caution was advised in interpreting the observed associations. Finally, while the reliability of current findings through various analyses was verified, such as E-value calculation, using age as time scale and sex as strata option, and Bonferroni correction, further studies are needed to comprehensively quantify these factors and their associations with hypertension.
The findings of this study underscored a significant link between the 10-year trajectories of LE8 and the risk of incident hypertension. Maintaining a high level of LE8 or improving LE8 status was linked to a decreased risk of hypertension. Notably, individuals with initially low LE8 levels could also experience a reduced risk of hypertension by enhancing their LE8 status, potentially even reversing their risk of hypertension. These results highlighted the importance of encouraging efforts aimed at optimizing and maintaining high CVH status among the public. Furthermore, they suggested that the LE8 could serve as a powerful tool for evaluating the risk of hypertension. By enhancing LE8 status, individuals and healthcare providers alike may effectively mitigate the risk of hypertension and promote CVH.
The datasets used and/or analyzed during the current study are available from the corresponding author upon reasonable request.
American Heart Association
Body mass index
Blood pressure
Blood lipids
Cardiovascular health
Confidence interval
China Health and Retirement Longitudinal Study
Fasting blood glucose
Hazard ratio
High-density lipoprotein cholesterol
high-sensitivity C-reactive protein
Life’s Simple 7
Life’s Essential 8
Physical activity
Standard deviation
Total cholesterol
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This study appreciates the participants and study staff/researchers involved in the Kailuan study.
Jiwen Zhong and Jinguo Jiang contributed equally to this work.
Department of Critical Care Medicine, Zhuhai People’s Hospital (Zhuhai Clinical Medical College of Jinan University), Zhuhai, 51900, Guangdong, China
Jiwen Zhong, Ruijun Huang & Wei Zheng
Department of Clinical Epidemiology, Shengjing Hospital of China Medical University, China Medical University, Shenyang, Liaoning, China
Jinguo Jiang
Department of Cardiology, Renmin Hospital of Wuhan University, 430060, Wuchang, Wuhan, China
Hubei Key Laboratory of Cardiology, 430060, Wuchang, Wuhan, China
Department of Epidemiology and Biostatistics, School of Public Health, Peking University, 38# Xueyuan Road, Haidian District, 100191, Beijing, China
Yang Liu & Xueying Qin
Key Laboratory of Epidemiology of Major Diseases (Peking University), Ministry of Education, Beijing, China
Xueying Qin
Department of Cardiology, Kailuan General Hospital, No.57 Xinhua East Road, 063000, Tangshan, Hebei Province, China
Shouling Wu & Shuohua Chen
Hypertension Center, State Key Laboratory of Cardiovascular Disease of China, National Center for Cardiovascular Diseases of China, Fuwai Hospital, Chinese Academy of Medical Sciences and Peking Union Medical College, 100037, Beijing, China
Department of Cardiology, Shenzhen Cardiovascular Minimally Invasive Medical Engineering Technology Research and Development Center, The Second Clinical Medical College, The First Affiliated Hospital, Shenzhen People’s Hospital, Jinan University, Southern University of Science and Technology), 518020, Shenzhen, Guangdong, China
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J.Z. and J.J.: Conceptualization, Software, Formal analysis, Writing - original draft, Writing - review & editing. L.G., Y.L., S.W., X.P., S.C., X.Q., and S.D.: Writing - review & editing., R.H., and W.Z.: Conceptualization, Review & editing, Supervision, Resources. J.Z. and J.J.: contributed equally as co-first authors. R.H. and W.Z.: contributed equally as co-corresponding authors. All authors read and approved the final manuscript.
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Zhong, J., Jiang, J., Guo, L. et al. 10-year trajectory of Life’s Essential 8 and incident hypertension: a community-based cohort study. Lipids Health Dis 23 , 278 (2024). https://doi.org/10.1186/s12944-024-02257-z
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Acute-phase reactions (APRs) are common among people treated for the first time with zoledronate (ZOL). The current view is that both the APRs caused by ZOL and its efficacy are related to the mevalonic acid pathway. However, the relationship between APRs and ZOL efficacy remains unclear.
This was a prospective observational cohort study involving postmenopausal women with osteoporosis in Shanghai, China, for 1 year. A total of 108 patients with an average age of 67.4 ± 5.8 years were treated with 5 mg intravenous ZOL for the first time. Data on demographic characteristics, APRs, blood counts, bone turnover markers, including C-telopeptide collagen crosslinks (CTX) and N-terminal propeptide of type 1 collagen (PINP), and bone mineral density (BMD) were collected.
(1) The results did not reveal a relationship between APRs and changes in bone turnover markers and BMD but showed that changes in body temperature (T) within 3 days after administration were positively correlated with changes in the BMD of the LS at Month 12 (β = 0.279 P = 0.034). (2) This effect was mediated mainly by changes in serum CTX (b = 0.046, 95% CI [0.0010–0.0091]). (3) The ROC curve revealed that when T increased by 1.95 °C, the sensitivity and specificity of identifying clinically important changes in LS BMD after 1 year were optimized.
In this study, we tested the hypothesis that people with elevated body T after initial ZOL treatment had greater improvements in BMD and better outcomes.
NCT, NCT03158246. Registered 18/05/2017.
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Osteoporosis has become a major public health problem with impacts on both quality of life and quantity of life. In China, the prevalence of osteoporosis among adults aged 40 years or older is 5.0% among men and 20.6% among women [ 1 ]. Bisphosphonates (BPSs) are the most commonly used therapy for osteoporosis [ 2 ].
Zoledronic acid (ZA), a third-generation nitrogen-containing BP, represents the most potent inhibitor of bone resorption and is long-acting in bone marrow, clearly improving the bone mineral density (BMD) and reducing the risk of vertebral, nonvertebral and hip fractures [ 3 , 4 ]. ZA has been recommended in guidelines as a first-line treatment for extremely high fracture risk [ 5 , 6 ]. Currently, ZOL is widely used in the clinic because of its significant efficacy, convenient annual administration, good patient compliance and few gastrointestinal reactions [ 7 ].
As a unique reaction after the administration of nitrogenous bisphosphonate (NBPS), acute-phase reactions (APRs) after the infusion of 5 mg of zoledronic acid for the first time are common. Fever and musculoskeletal pain occurred 24 to 36 h after NBP administration. Fever was noted to be associated with a decline in circulating lymphocyte number and increases in circulating γδ-T cells, IL-6 and tumour necrosis factor-alpha (TNF-a). These reactions are collectively referred as the acute phase response.
The main mechanism of ZOL is to inhibit the activity of farnesyl pyrophosphate (FPP) synthetase, the key enzyme of the mevalonate pathway, which can selectively suppress osteoclastic bone resorption. This mechanism may also be the basis for the occurrence of APRs [ 8 ]. ZOL inhibits FPP synthetase, resulting in the accumulation of isopentenyl diphosphate (IPP) and dimethylallyl diphosphate (DMAPP), the upstream intermediates of FPP synthetase of this metabolic pathway in monocytes, and then activates γδ- T cells via the release of proinflammatory cytokines, such as interleukin-6, interferon-γ and tumour necrosis factor β [ 9 , 10 , 11 ]. Therefore, APRs are more significant after ZOL treatment in those who are more sensitive to the mevalonate pathway. Consequently, both APRs caused by ZOL and its efficacy are likely related to the mevalonic acid pathway.
Approximately 30-54.9% of patients treated with a once yearly intravenous infusion of ZOL experienced an ARP, characterized by a transient flu-like syndrome with fever, chills, flushes, fatigue, malaise, and musculoskeletal and gastrointestinal symptoms following the first infusion [ 12 , 13 ]. Therefore, we designed the present study, which lasted for 1 year after the first ZOL treatment infusion. In this study, we explored the relationships between APR and changes in bone turnover markers and bone mineral density after the first application of ZOL in postmenopausal women with osteoporosis to assess whether the therapeutic effect could be predicted by the occurrence of APR after the first ZOL administration.
This was a prospective observational cohort study involving postmenopausal women with osteoporosis in Shanghai, China, for 1 year. The study was registered at ClinicalTrials.gov via the Protocol Registration and Results System (PRS) on 18/05/2017 (ClinicalTrials. gov ID: NCT03158246, First posted date 18/05/2017) by Cttq. The Formula n = 2[(Z α +Z β )σ/δ] 2 was used to calculate the sample size. We calculated the sample size that was necessary to achieve a statistical power of 80% with a significance level of 0.05. Thus, α was set at a level of 0.05, whereas β was set at a level of 0.2. Therefore, Z α was 1.96 (2-sided testing), whereas Z β was 0.84. According to our previous research, σ, which was the standard deviation of the change in BMD between the two groups, was 0.022; δ, the difference in the mean BMD between the two groups, was 0.009. Therefore, the calculated sample size (n) was determined to be 95. Considering the loss to follow-up, the sample size was increased by 15%, resulting in a final sample size of 110. The flowchart of this study is shown in Fig. 1 . A total of 110 patients were ultimately enrolled. A total of 108 patients with an average age of 67.4 ± 5.8 years and a mean body mass index of 22.3 ± 3.2 kg/m 2 who were treated with 5 mg intravenous ZOL for the first time between September 2017 and March 2018 completed all the visits and were included in the analysis. Two subjects were excluded from the study. The first subject withdrew her informed consent because she moved to another city 2 months after enrolling in the study. The second subject passed away in a car accident 7 days after enrolment. In our final analysis, we chose not to incorporate the data from these two patients because of an acceptable loss to follow-up rate of 1.8% and because their withdrawals were unrelated to the examination of drug side effects and therapeutic effects. A large amount of missing data can severely undermine the validity of inference and conclusions of a study. It is generally believed that if the proportion of missing values is less than 5%, samples containing missing values can be deleted, which will not have a significant effect on the overall analysis of the dataset. Otherwise, multiple imputation, weighting and maximum likelihood-based methods can be used to deal with incomplete data. Intravenous ZOL was administered at a dose of 5 mg over 15 min, and calcium (600 mg/d) and vitamin D (400 IU/d) were given as basic supplements following the infusion. To ensure compliance with treatment, calcium and vitamin D supplements were uniformly given by the investigators. A diary card was provided to request that the subjects record medication usage and adverse effects. Telephone interviews were conducted on Days 1 − 3 and at Months 3 and 9. Adverse events and calcium and vitamin D supplement consumption of the subjects were recorded. Clinical visits were conducted on Day 14 and at Months 6 and 12. The subjects were required to return the remaining calcium and vitamin D supplements and empty medicine bottles. Compliance with treatment was judged on the basis of the remaining amount of supplements. New calcium and vitamin D supplements were subsequently distributed for the next clinical visit. Oral T (T) at baseline and every morning for 3 days and on Day 14 after the infusion was recorded. Baseline and Day 14 blood samples were used to measure complete blood counts, liver and kidney function, and bone turnover marker (BTM) levels. Moreover, BTMs were also measured at Months 6 and 12. The BMDs of the lumbar spine (L2–4) and left total hip were measured at baseline and at 6 and 12 months after infusion. The investigators were blinded to the treatment allocation. The technicians involved in the measurement of BMD and BTM were blinded to any discernible clinical information regarding the participants, employing a rigorous blinding process. The study protocol and procedures were approved by the ethics committee of the hospital (No: 20170068).
Flowchart of this study
Postmenopausal women aged from 46 to 80 years with osteoporosis eligible for intravenous ZOL were eligible for inclusion if they had a bone mineral density T score of -2.5 or less at either the lumbar spine, hip, or femoral neck; a T score between − 1.0 and − 2.5; fragility fracture of the proximal humerus, pelvis, or distal forearm; or low-trauma spine or hip fracture regardless of BMD. Moreover, they had to be capable of monitoring and recording basic data on their body T and clinical symptoms.
The exclusion criteria were as follows: (1) patients who had previously used bisphosphonates; (2) patients who had used parathyroid hormone; (3) patients who had undergone surgical treatment in the past three months; (4) patients with a total serum calcium level of less than 2.1 mmol/L (8.4 mg/dL) or a serum-free calcium level of less than 0.95 mmol/L (3.8 mg/dL) or untreated hypocalcaemia; (5) patients with a Cockcroft calculated creatinine clearance of less than 35 mL/min; and (6) patients who refused to take ZOL, those with causes of osteoporosis other than postmenopausal osteoporosis, and patients with other diseases with an unstable status. Informed consent was obtained from all individual participants included in the study.
Basic information collection and anthropometric measurements.
Medical history, including fracture history and intake of antiosteoporosis drugs, was assessed, and a physical examination was performed. Height was measured using a stadiometer to the nearest 0.01 m. Body weight was measured by applying a standard balance beam scale to the nearest 0.1 kg. Body mass index (BMI) was calculated as the body weight divided by the square of the height (kg/m 2 ).
The BMDs of the lumbar spine (L2–4) and left total hip were measured with a dual-energy X-ray absorptiometry densitometer (Hologic Delphi A; Hologic Inc., Methuen, MA, USA). The precision error in our laboratory was 0.8% for the lumbar spine, 1.05% for the femur neck and 0.97% for the total hip. The densitometer was standardized with a standard phantom before each measurement.
All blood samples were obtained in the morning after a 10 h overnight fast and were stored immediately at − 80 °C for subsequent assays. N-terminal propeptides of type 1 collagen (P1NP), C-telopeptide collagen crosslinks (CTXs) and 25OHD were measured via electrochemical luminescence (Roche Diagnostics, Boston, MA, USA), with intra- and interassay coefficients of variance (CVs) below 3.5% and 8.4% for CTX, below 2.6% and 4.1% for P1NP, and below 7.8% and 10.7% for 25OHD. Blood counts included white blood cell (WBC), neutrophil (N), lymphocyte (L), monocyte (M), eosinophil (E), basophil (B), red blood cell (RBC), and platelet (PLT) counts.
Previous analyses of APRs have been performed in several studies, and we used a similar definition for our study [ 14 ]. The definition utilized adverse events, which were reported as occurring within 3 days of the first administration of ZOL. The adverse events were categorized according to the Medical Dictionary for Regulatory Activities (MedDRA) version 9 [ 15 ]. Preferred terms meeting the definition of APR were grouped into five symptom clusters: fever; musculoskeletal events (e.g., pain and joint swelling); gastrointestinal events (e.g., abdominal pain, vomiting, and diarrhoea); eye inflammation; and other events (including fatigue, nasopharyngitis, and oedema). Oral T was checked in each patient at 06:00 at baseline and on Days 1 to 3 and 14 after ZOL administration. If a patient’s body temperature exceeded 37.3 °C, it was remeasured every 4 h. Standardized digital thermometers were utilized for all temperature measurements. Body temperature data were centralized and stored in a single location rather than being dispersed across multiple sites or investigators. Thresholds were based on prior study recommendations [ 16 ].
SPSS v23 (SPSS Inc., Chicago, IL, USA) was used to analyse the data. The Shapiro‒Wilk test was performed to determine whether variables were normally distributed. Continuous variables are expressed as means with standard deviations. Skewed data are presented as the median with the interquartile range (25–75%). Classification variables are expressed as percentages. Data that were not normally distributed (all blood counts) were log transformed to normality before being used in the statistical analyses. After normality, independence, and equal variance (homoscedasticity) of the variables were checked, two-sample t tests or Mann‒Whitney U tests were used to analyse differences between groups. After the independence of variables (noncollinearity), normality of the residuals, independence between observations, and equal variances were checked, multivariate linear regression models were constructed to analyse the correlation between dependent variables, such as BMD, and independent continuous variables, such as the change in temperature, PINP, CTX, and blood count indices. The SPSS PROCESS macro programme was used to test the mediating effect of CTX on the relationship between T and BMD. The analysis facilitated estimation of the indirect effect with a normal theory approach and a bootstrap approach to obtain confidence intervals [ 17 ]. In PROCESS, Model 4 with 10,000 interactions was established to determine the mediation of the regression models, CTX with T and BMD. The regression coefficient of T to CTX was denoted as path a, as shown in Fig. 2 . Second, the regression coefficient of T to BMD as the total effect was denoted path c. Third, the regression coefficients for both T and CTX to BMD as the direct effect were denoted paths b and c. Finally, the indirect effect was examined by 95% bootstrapped confidence intervals (CIs) using 10,000 bootstrapped samples. The receiver operating characteristic (ROC) curve and area under the curve (AUC) were used to estimate the utility, and the cut-off value, sensitivity and specificity were calculated simultaneously. Statistical significance was set at P < 0.05.
Mediating effect of change in CTX between change in T and change in BMD. CTX: cross-linked C-telopeptide of type I collagen; T: temperature; BMD: bone mineral density AUC: area under the curve; BMD: bone mineral density; LS: lumbar spine
The baseline and follow-up characteristics of the subjects are shown in Table 1 . The body T increased significantly within 3 days after the administration of ZOL and returned to the baseline level on Day 14. The BMD of the lumbar spine (LS) and femur neck (FN) increased significantly at 6 and 12 months after treatment. Compared with the baseline level, the BMD of the LS increased by 2.2% (Month 6) and 2.5% (Month 12), and the BMD of the FN increased by 1.7% (Month 6) and 2.1% (Month 12). In terms of bone metabolism turnover markers, the level of serum CTX decreased significantly after 2 weeks of treatment, with a decrease of 92.0%. The level of CTX increased slightly at 6 and 12 months after treatment, but it was still significantly lower than the baseline level, with decreases of 80.0% and 71.5%, respectively. P1NP did not change significantly 2 weeks after treatment but decreased at 6 and 12 months after treatment, with decrease rates of 62.9% (Month 6) and 56.0% (Month 12), respectively. The WBC count increased 2 weeks after treatment, the proportion of neutrophils increased substantially, and the percentages of lymphocytes and monocytes decreased.
As shown in Table S1 , 75.9% of the subjects had APRs within 3 days after treatment, of which fever and chills were the most common, accounting for 64.8% of the subjects, followed by musculoskeletal pain, accounting for 38.9% of the subjects, and gastrointestinal, ophthalmic and other symptoms accounted for 13.0%, 3.7% and 29.6% of the subjects, respectively. The subjects were categorized into two groups by APR occurrence: APR+ (i.e., with APR) and APR- (i.e., without APR) (Table 2 ). The results revealed no significant difference between the two groups in BMD changes in the LS, FN or total hip (TH) at Months 6 and 12 (Table 2 ).
Multivariate linear regression models were constructed to analyse the correlations between changes in body T within 3 days and 25OHD at baseline, and changes in BTM (baseline vs. 12 months) and changes in BMD (baseline vs. 12 months) are shown in Table 3 . The results showed that after adjustment for age and BMI, changes in T (baseline vs. 3 days) were positively correlated with changes in LS BMD (baseline vs. 12 months) (β = 0.270 P = 0.034), and changes in CTX (baseline vs. 12 months) were negatively correlated with LS BMD (baseline vs. 12 months) (β=-0.486 P = 0.000). The change in PINP (baseline vs. 12 months) was negatively correlated with the change in LS BMD (baseline vs. 12 months) (β=-0.409 P = 0.001). The 25OHD at baseline was not correlated with the change in BMD. None of the factors were correlated with changes in FN BMD or TH BMD (baseline vs. 12 months).
Table 4 ; Fig. 2 illustrate the results of the bootstrapped mediation analysis to determine whether CTX mediated the relationship between the change in T BMD and the change in LS BMD. With respect to the total effect, the change in T was positively linked to the change in LS BMD (b = 0.123, P = 0.00371). Furthermore, the change in T was negatively correlated with the change in the serum CTX concentration. The change in serum CTX was negatively correlated with the change in BMD. Moreover, CTX had a significant mediating effect on the relationship between the change in T and the change in the BMD of LS patients (b = 0.046, 95% CI [0.0010–0.0091]). However, there was no direct effect between the change in T and the change in BMD. The results suggested that the influence of the change in T on the change in the BMD of the LS was mainly mediated by the change in CTX.
According to the clinical consensus, a change in BMD by more than 2.77 times the precision error is defined as the least significant change (LSC) [ 18 ]. In this study, the precision errors were 0.80% for the spine, 1.05% for the femur neck and 0.97% for the total hip; thus, the LSCs were 2.2% for the spine, 2.90% for the femur neck and 2.67% for the total hip. Therefore, a change in LS BMD of more than 2.2% at 12 months was defined as a clinically significant change. The ROC curve was used to observe the sensitivity and specificity of the change in T to judge the clinically significant change in LS BMD. The results showed that when T increased by 1.95 °C, the sensitivity and specificity of judging the clinical significance change in LS BMD were optimized. The sensitivity and specificity were 68.2% and 78.1%, respectively, and the area under the curve was 0.724 (Fig. 3 ).
Change in T to predict the increase in BMD in LS
In this study, we investigated the hypothesis that the treatment effect in patients with APRs after the first administration of ZOL was better than that in patients without APRs. However, the results did not reveal a relationship between the APRs and changes in bone turnover markers and BMD but revealed that changes in T within 3 days after administration were positively correlated with changes in the BMD of the LS at month 12 (β = 0.270 P = 0.034). This effect was mediated mainly by changes in the serum CTX concentration (b = 0.046, 95% CI [0.0010–0.0091]). Moreover, the ROC curve revealed that when T increased by 1.95 °C, the sensitivity and specificity of judging the clinical importance of the change in LS BMD after 1 year were optimized.
APRs are common among people who are treated for the first time with ZOL. The incidence varies from 30–54.9% [12–13,19−20] . APRs generally appear within 3 days following the first infusion, and common symptoms include fever, musculoskeletal pain, and gastrointestinal symptoms [ 19 ]. APRs are more common among younger subjects, those using nonsteroidal anti-inflammatory drugs (NSAIDs), and non-Japanese Asians, whereas they are less common in smokers, patients with diabetes, previous oral bisphosphonate users and Latin Americans [ 20 ]. Currently, research on APRs has focused mostly on drug safety. However, according to previous studies, the mevalonate pathway that causes APRs is also related to the efficacy of the medication. Therefore, we hypothesized that people with APRs after the first application of ZOL would experience better efficacy after treatment. At present, only 3 studies have been conducted to examine the relationship between APRs and treatment efficacy, and the results were inconsistent [ 21 , 22 , 19 ]. Subanalyses of the phase III ZONE study conducted in Japan demonstrated that patients with APRs presented significantly greater increases in total hip BMD at 6 and 12 months and greater suppression of BTMs compared with patients without APRs [ 22 ]. Another study conducted by Lu suggested a potential association between APRs occurrence and decreased refracture risk in patients with osteoporotic fractures and in patients undergoing orthopaedic surgery [ 19 ]. The HORIZON Pivotal Fracture Trial showed no significant difference in treatment-related changes in BMD with or without APRs; however, subjects starting ZOL who experienced APRs had a greater reduction in vertebral fracture risk [ 21 ]. In addition to racial differences, an important reason for this inconsistency lies in the choice of research methods. APRs are composed of a group of symptoms, including fever, musculoskeletal pain, gastrointestinal symptoms, and eye symptoms. In fact, most of these symptoms are self-reported by patients and are difficult to quantify and compare objectively, resulting in deviations in the results. Fever is the most common symptom in APRs. In studies in Japan and the United States, 75.7% and 47.5% of people with APRs, respectively, had fever symptoms [ 21 , 22 ]. Therefore, in this study, body T, the most common and easily quantifiable indicator of APRs, was selected as the observation indicator to analyse its possible correlation with the efficacy of ZOL after treatment, including changes in BTM and BMD.
Our results revealed that when grouped by APRs, there were no significant differences in changes in BMD in the LS, FN, or TH between the two groups at Months 6 and 12, which is consistent with the results reported in the HORIZON-PFT study. However, when we used the change in body T within 3 days after the first infusion of ZOL as an observation indicator, the results showed that there wasrevealed a significant positive correlation between the changes in T (baseline vs. 3 days) and LS BMD (baseline vs. 12 months) (β = 0.270 P = 0.034). Moreover, further mediating effect analysis revealed that the influence of the change in T on the change in the BMD of LS patients was mediated mainly by the change in the serum CTX level (b = 0.046, 95% CI [0.0010–0.0091]).
Bone continuously undergoes modelling and remodelling to maintain bone metabolism and structural completion. This process of self-renewal, in which osteoclasts constantly absorb old bone and osteoblasts constantly form new bone, is known as bone turnover. Bone turnover markers (BTMs) are metabolic products or enzymes produced during bone turnover processes and are classified as bone formation or bone resorption markers. Serum CTX is a sensitive marker recommended by the IOF to reflect bone resorption [ 23 ]. Our results revealed a negative correlation between the changes in serum CTX levels at Week 2, Month 6, and Month 12 after infusion of ZOL and the changes in BMD in the LS and TH at Month 12. Although our study did not evaluate the occurrence of fractures because it was a one-year clinical trial, according to the reported studies, compared with those with the lowest quartile level of serum CTX, the risk of vertebral fracture increased by 1.4–2.2 times, and the risk of nonvertebral fracture increased by 1.8–2.5 times in the subjects with the highest quartile of serum CTX independent of BMD [ 24 ]. The greater the decrease in serum level of CTX, the greater is the decrease in fracture risk [ 25 ].
Notably, our results suggested that the impact of changes in body T on serum CTX after infusion of ZOL may be related to monocytes. The changes in T within 3 days after treatment were negatively correlated with the changes in serum CTX levels at Month 6 and Month 12. Moreover, the changes in T were also negatively correlated with the changes in monocytes in the peripheral blood at Week 2 (Table S2 ). Next, we subclassified the subjects into four groups according to the degree of monocyte reduction after treatment, and the changes in serum CTX were compared between the lowest quartile and highest quartile groups. The results revealed that the decrease in serum CTX was greater in subjects in the highest quartile group compared with those in the lowest quartile group, and the decrease in CTX at Month 12 was significantly different (− 78.2% vs. − 57.8%, P < 0.05) (Table S3 ).
Currently, both peripheral monocytes and γδ T cells are believed to be rapidly activated after treatment with ZOL, which ultimately affects the clinical severity of APR [ 20 , 26 , 27 ]. ZOL is taken up by monocytes, which thereby acquire the ability to ‘‘present’’ isoprenoid metabolites or related structures to γδ T cells [ 28 ]. Through mutual crosstalk, both monocytes and γδ T cells undergo a series of activation and differentiation steps that ultimately determine the severity of the APRs and replicate similar events to those that occur in acute infection [ 9 ]. Notably, the effects on circulating monocytes appear to be transient, lasting only 3 days; subsequently, rapid inactivation and extravasation occur [ 20 ]. Therefore, in our study, we observed a significant decrease in the number of monocytes at week 2 after treatment. On the one hand, the number of decreased monocytes reflects the severity of APRs in the body. On the other hand, it also reflects the sensitivity of osteoclasts to ZOL. Osteoclasts differentiate from monocyte-derived macrophage precursors and are responsible for bone absorption. Therefore, changes in the number of serum mononuclear cells after treatment can partially indicate the sensitivity of osteoclasts to ZOL. As shown in our study, the subjects with greater reductions in monocyte counts also presented greater decreases in serum CTX levels. The level of CTX reflects the activity of osteoclasts and the level of bone resorption.
Special attention should be given to a recent study conducted by Lu, which focused on elderly osteoporotic fracture patients undergoing orthopaedic surgery [ 19 ]. In that study, the long-term outcomes of APRs after initial ZOL administration were investigated. The results demonstrated that patients who experienced APR had a 73% lower rate of refracture than did those who did not, which indicated that patients with APRs experienced better drug efficacy. Our research yielded similar findings, showing that the increase in body T within 3 days after initial ZOL was related to changes in BTM and an increase in BMD after one year. However, surprisingly, Lu’s research also revealed that patients who had an APR had a 97% higher risk of mortality than patients who did not. In Lu’s study, all the subjects presented newly identified hip/morphological vertebral osteoporotic fractures and had undergone orthopaedic surgery in the hospital. Most of the subjects had received ZOL treatment for 7 − 14 days after the orthopaedic operation. In fact, acute fracture events and surgery comprise a special period for patients with osteoporosis. Research has shown that trauma itself can exacerbate the occurrence of ARPs [ 29 ]. Compared with those after nonsurgical interventions, the odds ratios of experiencing APRs after minimally invasive or open surgery were 3.54 and 5.71, respectively. Therefore, the results of Lu’s study can only represent the impact of ARPs on mortality and refracture after ZOL treatment of hospitalized osteoporotic fracture patients. However, the subjects in our study were recruited from the outpatient service, and none of them had undergone surgical treatment in the past three months. Therefore, our results are more universally applicable to patients with osteoporosis.
Both the APRs and efficacy of ZOL were related to the mevalonate pathway. Our results suggested that those patients who are sensitive to ZOL were more likely to have an increase in body T, a greater decrease in serum CTX and a greater increase in BMD after the first dose. A three-year HORIZEN-PFT study has shown that people with APRs are less likely to suffer vertebral fractures [ 21 ]. In our study, the degree of increase in body T after initial medication might indicate the sensitivity of patients to ZOL and potentially predict the treatment efficacy to some extent. Our study further revealed that an increase in body T of 1.95 °C or more within 3 days after the first administration of ZOL predicted a clinically significant increase in lumbar bone density at 1 year after the first administration of ZOL.
There are several limitations to our study. (1) While this study was conducted as a cohort study, it is worth noting that the follow-up period was limited to one year. The current view is that there is a significant change in BTM after 3 months of treatment with ZOL. A significant change in BMD was observed after 6 months of treatment with ZOL. Moreover, the fastest increase in BMD can be observed after the first year of treatment with ZOL [ 30 , 31 ]. Therefore, osteoporosis efficacy, including changes in BTM and BMD, can be evaluated over one year of follow-up. However, the relatively short duration of the follow-up period might account for the observation limited to changes in BTM and BMD, with no occurrences of fractures recorded within our study cohort. Three existing studies have investigated the long-term effects of zoledronic acid in patients with and without APR. One of them is the HORIZON PFT study, which lasted for 3 years and focused on postmenopausal women. BMD and fracture occurrence were observed [ 21 ]. The results suggest that women initiating ZOL who experience an APR will have a greater reduction in vertebral fracture risk with ZOL. However, there was no significant difference in treatment-related changes in BMD. Another study conducted in Japanese patients treated with a once yearly intravenous infusion of ZOL 5 mg for 2 years investigated the relationship between APR and efficacy [ 22 ]. Patients with APRs presented significantly greater increases in total hip BMD at 6 and 12 months and greater reductions in BTMs than patients without APRs. However, fracture risk was not assessed in this study. The other is a retrospective study involving elderly osteoporotic fracture patients who underwent orthopaedic surgery, which has been discussed above [ 19 ]. (2) Despite meeting the requirements established by our power analysis, the sample size was still relatively small. (3) Given the higher prevalence of osteoporosis among postmenopausal women, the study participants in this exploratory study were limited to postmenopausal women residing in Shanghai. The results of this study present evidence that individuals exhibiting elevated body T levels following initial ZOL treatment may experience greater improvements in BMD. (4) In this study, certain external variables, such as lifestyle and nutritional status, were not collected, which might compromise the robustness of the findings. To gain a more comprehensive understanding of the long-term efficacy, safety, and sustained effects of ZOL, further studies with longer follow-up durations are needed. In future studies with longer follow-up durations (e.g., longer than 2 years), the change in occurrence of fractures can be better evaluated. In addition, in further research, external variables such as lifestyle, exercise and dairy intake will be collected to increase the validity of the results.
In summary, the results of this study indicated for the first time that (1) there was no significant correlation between the occurrence of ARPs within 3 days of the first administration of ZOL and the change in BMD after 1 year, whereas the increase in body T positively correlated with the increase in BMD, suggesting that the increase in body T predicted better treatment. (2) The correlation between the increase in body T and increase in BMD was mediated by the change in serum CTX. (3) An increase in body T of 1.95 °C or above predicted that the BMD in the LS would increase significantly 1 year after the first administration of ZOL.
In this study, we tested the hypothesis that people with elevated body T after initial ZOL treatment had greater improvements in BMD and better outcomes. The establishment of this hypothesis will be conducive to the prediction of the clinical curative effect and guide doctors in timely intensive treatment for people with poor curative effects. It is widely recommended in guidelines that the measurement of main serum bone turnover biomarkers, namely, CTX and PINP, should be performed over a time period ranging from 3 − 6 months posttreatment for assessment of drug efficacy. However, the use of a body temperature monitor allows for a timelier, cost-effective, and convenient assessment in comparison, particularly in regions where access to these laboratory tests may be limited. More importantly, since APRs are very common in patients with initial ZOL treatment, the establishment of this hypothesis will help alleviate patients’ fear and anxiety surrounding APRs, improve patients’ confidence and compliance with antiosteoporosis therapy, and increase humanistic care for patients.
The datasets used and/or analysed during the current study are available from the corresponding author on reasonable request.
Area under the curve
Acute-phase reactions
Bone turnover marker
Bone mineral density
Body mass index
Bisphosphonates
Confidence intervals
C-telopeptide collagen crosslinks
Coefficients of variance
Dimethylallyl diphosphate
Farnesyl pyrophosphate
Isopentenyl diphosphate
Lumbar spine
Least significant change
Medical Dictionary for Regulatory Activities
Mevalonate kinase
Nitrogenous bisphosphonate
Nonsteroidal anti-inflammatory drugs
N-terminal propeptide of type 1 collagen
Phosphomevalonate kinase
Red blood cell
Operating characteristic curve
Temperature
White blood cell
Zoledronate
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This study was supported by National Key R&D Program of China (2018YFC2000205), Shanghai Municipal Health Commission (20214Y0519), Bethune Foundation Project (G-X-2019-1107-2), Bethune Foundation Project (GX2021C04), Shanghai Science and Technology Commission Science and Technology Innovation Action Plan Industry University Research Medical Cooperation Project (17DZ1920206). The funding body had no role in the design of the study and collection, analysis, interpretation of data or in writing the manuscript.
Yanping Du and Weijia Yu contributed equally to this work and should be considered co-first author.
Department of Osteoporosis and Bone Disease, Research Section of Geriatric Metabolic Bone Disease, Huadong Hospital Affiliated to Fudan University, Shanghai Geriatric Institute, Shanghai, China
Yanping Du, Weijia Yu, Wenjing Tang, Minmin Chen, Huilin Li & Qun Cheng
Research Section of Geriatric Metabolic Bone Disease, Shanghai Geriatric Institute, Shanghai, China
Department of Traditional Chinese Medicine, Huadong Hospital Affiliated to Fudan University, Shanghai, China
Haixin Gou & Yiming Lei
Department of Massage, Huadong Hospital Affiliated to Fudan University, Shanghai, China
Tongkai Zhang
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QC has contributed to the conception and design of this study. YD, WY performed statistical analysis and wrote the first draft of the manuscript. QC, YD, WY, WT, MC, HL revised the manuscript and provided the meritorious support. YD, WY, HG, YL, TZ, WT, MC, HL participated in recruitment and examinations of patients, they also contributed to the acquisition and analysis of data; HG, YL, TZ performed laboratory analysis. All authors interpreted data. All authors approved the final version of the manuscript.
Correspondence to Qun Cheng .
Ethics approval and consent to participate.
All procedures performed in studies involving human participants were in accordance with the ethical standards of the institutional and national research committee (The Ethics Committee of The Huadong Hospital, No: 20170068) and with the 1964 Helsinki declaration and its later amendments or comparable ethical standards. Written informed consent was obtained from all individual participants prior to enrollment. Ethics approval was obtained from Ethics committee of The Huadong Hospital.
The authors declare no competing interests.
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Du, Y., Yu, W., Gou, H. et al. Change in body temperature, not acute-phase reaction, predict anti-Osteoporosis efficacy after the first administration of Zoledronic acid: a prospective observational cohort study. BMC Musculoskelet Disord 25 , 694 (2024). https://doi.org/10.1186/s12891-024-07781-8
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DOI : https://doi.org/10.1186/s12891-024-07781-8
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See Section III. 3. Additional Information on Eligibility .
This Notice of Funding Opportunity (NOFO) encourages grant applications from investigators interested in conducting basic, mechanistic research into the biological/genetic causes of cancer health disparities. These research project grants (R01) will support innovative studies designed to investigate biological/genetic bases of cancer health disparities, such as (1) mechanistic studies of biological factors associated with cancer health disparities, including those related to basic research in cancer biology or cancer prevention strategies, (2) the development and testing of new methodologies and models, and (3) secondary data analyses. This NOFO is also designed to aid and facilitate the growth of a nationwide cohort of scientists with a high level of basic research expertise in cancer health disparities research who can expand available resources and tools, such as biospecimens, patient derived models, and methods that are necessary to conduct basic research in cancer health disparities.
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There are several options available to submit your application through Grants.gov to NIH and Department of Health and Human Services partners. You must use one of these submission options to access the application forms for this opportunity.
Section i. notice of funding opportunity description.
This notice of funding opportunity (NOFO) encourages grant applications from investigators interested in conducting basic, mechanistic research into the biological/genetic causes of cancer health disparities. These research project grants (R01) will support innovative studies designed to investigate biological/genetic bases of cancer disparities, such as (1) mechanistic studies of biological factors associated with cancer disparities, including those related to basic research in cancer biology or cancer prevention strategies, (2) the development and testing of new methodologies and models, and (3) secondary data analyses. This NOFO is also designed to aid and facilitate the growth of a nationwide cohort of scientists with a high level of basic research expertise in cancer health disparities research who can expand available resources and tools, such as biospecimens, patient-derived models, and methods that are necessary to conduct basic research in cancer health disparities.
Please note that this NOFO will be reissued on the simplified review framework (SRF) template in 2025 to allow for resubmissions and renewals. See Simplified Review Framework for NIH Research Project Grant Applications.
In the United States, several racial/ethnic populations demonstrate increased incidence and/or more aggressive disease for specific cancer types. For example, African American males have higher rates of prostate and lung cancer, compared to their Caucasian-American counterparts, and both males and females exhibit a higher incidence of colorectal cancer and multiple myeloma. Hispanic/Latino individuals have the highest rates of cervical cancer and pediatric acute lymphoblastic leukemia (ALL) and one of the poorest survival rates for ALL. Similarly, Asians and Pacific Islanders have the highest incidence rates for liver and stomach cancer while American Indians and Alaska Natives have both the highest incidence and the highest mortality rates in kidney and renal pelvis cancer.
The causes of these cancer health disparities are multifactorial, including barriers in access to healthcare, cultural barriers, environmental disadvantage, differences in diet and lifestyle, ancestry-related risk factors, persistent co-morbidities, and chronic stress exposure due to discrimination and social isolation. An increasing number of studies demonstrate that even when socioeconomic and access to care factors are accounted for, incidence and mortality gaps persist between racial/ethnic populations for some cancer types, which suggests a role for biological contributors. Such studies have included identification of ancestry-related differences in DNA, RNA, and/or protein expression that are associated with cancer risk and/or progression. Other studies have shown the presence of differential tumor microenvironment components among diverse racial/ethnic populations indicating a potential role for immunity and inflammation in contributing to cancer health disparities.
These complex biological factors may enhance understanding of the differences observed in cancer incidence, prevalence, morbidity, and mortality rates among underrepresented populations. The NCI encourages investigations of such biological factors to increase our understanding of the mechanisms that play a role in cancer health disparities.
The goal of this NOFO is to stimulate interest in the characterization and functional analysis of biological factors associated with cancer health disparities and to provide funding opportunities in this area. Applications should focus on basic cancer research, consistent with the research interests of the NCI's Division of Cancer Biology (DCB) , Division of Cancer Prevention (DCP) , and Center for Cancer Health Equity (CCHE):
The DCB supports research on the discovery and characterization of basic pathways and mechanisms that regulate the development of a pre-malignant state, initiation of cellular transformation and cancer cell progression, formation of the tumor microenvironment, metastasis, and host responses to cancer, including immunologic or metabolic responses.
The CCHE supports cancer health disparity research focused on basic, hypothesis-driven studies that explicitly address the unequal burden of cancer amongst racial/ethnic minorities or other underserved populations across the cancer continuum (prevention, early detection, diagnosis, treatment, and survivorship).
The DCP supports research that will generate new information about molecular processes that are susceptible to intervention throughout the cancer continuum until invasive cancer and underlying mechanisms of cancer and its sequelae (i.e., mechanistic studies on the prevention or treatment of acute and chronic symptoms and morbidities related to cancer and its treatment), developing effective cancer screening and prevention strategies, discovering early detection biomarkers, and pinpointing mechanistically targeted nutrients in cancer prevention.
This NOFO encourages basic research projects that will develop and test new methodologies and new research technologies focused on specific topics in cancer health disparities. The availability of annotated clinical samples as well as enabling technologies (genomics/epigenomics, proteomics, metabolomics, single-cell analysis, imaging) make it feasible to study biological factors that contribute to cancer health disparities among different racial/ethnic populations.
Research projects must propose to investigate the interplay of race/ethnicity and/or other social determinants with cancer biology to mechanistically explain an unequal burden of cancer among populations. As such, proposed studies are encouraged to use biospecimens, patient-derived models, and/or data sets derived from different racial/ethnic and/or underserved groups. Studies investigating age and/or gender disparities, in the absence of race/ethnicity variables, are not solicited. Research projects using a comparative research design between at least two populations are encouraged, in which one or more is underserved.
Projects that are strictly hypothesis-generating, exploratory, and correlative studies are discouraged. As this NOFO is focused upon basic research, immediate clinically translational potential of the proposed project is NOT a requirement for the proposed projects.
Research topics of interest include but are not limited to :
The following types of studies are not responsive to this NOFO- applications proposing such studies will be considered non-responsive and will not be reviewed or considered for funding:
See Section VIII. Other Information for award authorities and regulations.
Grant: A financial assistance mechanism providing money, property, or both to an eligible entity to carry out an approved project or activity.
The OER Glossary and the How to Apply - Application Guide provide details on these application types. Only those application types listed here are allowed for this NOFO.
Not Allowed: Only accepting applications that do not propose clinical trials.
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The scope of the proposed project should determine the project period. The maximum project period is 5 years.
NIH grants policies as described in the NIH Grants Policy Statement will apply to the applications submitted and awards made from this NOFO.
1. eligible applicants eligible organizations higher education institutions public/state controlled institutions of higher education private institutions of higher education the following types of higher education institutions are always encouraged to apply for nih support as public or private institutions of higher education: hispanic-serving institutions historically black colleges and universities (hbcus) tribally controlled colleges and universities (tccus) alaska native and native hawaiian serving institutions asian american native american pacific islander serving institutions (aanapisis) nonprofits other than institutions of higher education nonprofits with 501(c)(3) irs status (other than institutions of higher education) nonprofits without 501(c)(3) irs status (other than institutions of higher education) for-profit organizations small businesses for-profit organizations (other than small businesses) local governments state governments county governments city or township governments special district governments indian/native american tribal governments (federally recognized) indian/native american tribal governments (other than federally recognized) federal governments eligible agencies of the federal government u.s. territory or possession other independent school districts public housing authorities/indian housing authorities native american tribal organizations (other than federally recognized tribal governments) faith-based or community-based organizations regional organizations non-domestic (non-u.s.) entities (foreign organizations) foreign organizations non-domestic (non-u.s.) entities (foreign organizations) are eligible to apply. non-domestic (non-u.s.) components of u.s. organizations are eligible to apply. foreign components, as defined in the nih grants policy statement , are allowed. required registrations applicant organizations applicant organizations must complete and maintain the following registrations as described in the how to apply - application guide to be eligible to apply for or receive an award. all registrations must be completed prior to the application being submitted. registration can take 6 weeks or more, so applicants should begin the registration process as soon as possible. failure to complete registrations in advance of a due date is not a valid reason for a late submission, please reference nih grants policy statement section 2.3.9.2 electronically submitted applications for additional information system for award management (sam) – applicants must complete and maintain an active registration, which requires renewal at least annually . the renewal process may require as much time as the initial registration. sam registration includes the assignment of a commercial and government entity (cage) code for domestic organizations which have not already been assigned a cage code. nato commercial and government entity (ncage) code – foreign organizations must obtain an ncage code (in lieu of a cage code) in order to register in sam. unique entity identifier (uei) - a uei is issued as part of the sam.gov registration process. the same uei must be used for all registrations, as well as on the grant application. era commons - once the unique organization identifier is established, organizations can register with era commons in tandem with completing their grants.gov registrations; all registrations must be in place by time of submission. era commons requires organizations to identify at least one signing official (so) and at least one program director/principal investigator (pd/pi) account in order to submit an application. grants.gov – applicants must have an active sam registration in order to complete the grants.gov registration. program directors/principal investigators (pd(s)/pi(s)) all pd(s)/pi(s) must have an era commons account. pd(s)/pi(s) should work with their organizational officials to either create a new account or to affiliate their existing account with the applicant organization in era commons. if the pd/pi is also the organizational signing official, they must have two distinct era commons accounts, one for each role. obtaining an era commons account can take up to 2 weeks. eligible individuals (program director/principal investigator) any individual(s) with the skills, knowledge, and resources necessary to carry out the proposed research as the program director(s)/principal investigator(s) (pd(s)/pi(s)) is invited to work with their organization to develop an application for support. individuals from diverse backgrounds, including underrepresented racial and ethnic groups, individuals with disabilities, and women are always encouraged to apply for nih support. see, reminder: notice of nih's encouragement of applications supporting individuals from underrepresented ethnic and racial groups as well as individuals with disabilities , not-od-22-019 , and notice of nih's interest in diversity, not-od-20-031 . for institutions/organizations proposing multiple pds/pis, visit the multiple program director/principal investigator policy and submission details in the senior/key person profile (expanded) component of the how to apply - application guide . 2. cost sharing.
This NOFO does not require cost sharing as defined in the NIH Grants Policy Statement NIH Grants Policy Statement Section 1.2 Definition of Terms.
Number of Applications
Applicant organizations may submit more than one application, provided that each application is scientifically distinct.
The NIH will not accept duplicate or highly overlapping applications under review at the same time, per NIH Grants Policy Statement Section 2.3.7.4 Submission of Resubmission Application . This means that the NIH will not accept:
1. requesting an application package.
The application forms package specific to this opportunity must be accessed through ASSIST, Grants.gov Workspace or an institutional system-to-system solution. Links to apply using ASSIST or Grants.gov Workspace are available in Part 1 of this NOFO. See your administrative office for instructions if you plan to use an institutional system-to-system solution.
It is critical that applicants follow the instructions in the Research (R) Instructions in the How to Apply - Application Guide except where instructed in this notice of funding opportunity to do otherwise. Conformance to the requirements in the How to Apply - Application Guide is required and strictly enforced. Applications that are out of compliance with these instructions may be delayed or not accepted for review.
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The following section supplements the instructions found in the How to Apply – Application Guide and should be used for preparing an application to this NOFO.
All instructions in the How to Apply - Application Guide must be followed.
Sf424(r&r) other project information, sf424(r&r) senior/key person profile, r&r or modular budget, r&r subaward budget, phs 398 cover page supplement, phs 398 research plan.
All instructions in the How to Apply - Application Guide must be followed, with the following additional instructions:
Resource Sharing Plan : Individuals are required to comply with the instructions for the Resource Sharing Plans as provided in the How to Apply - Application Guide .
Other Plan(s):
All instructions in the How to Apply - Application Guide must be followed, with the following additional instructions:
Appendix: Only limited Appendix materials are allowed. Follow all instructions for the Appendix as described in the How to Apply - Application Guide .
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If you answered Yes to the question Are Human Subjects Involved? on the R&R Other Project Information form, you must include at least one human subjects study record using the Study Record: PHS Human Subjects and Clinical Trials Information form or Delayed Onset Study record.
Study Record: PHS Human Subjects and Clinical Trials Information
Delayed Onset Study
Note: Delayed onset does NOT apply to a study that can be described but will not start immediately (i.e., delayed start). All instructions in the How to Apply - Application Guide must be followed.
Foreign organizations.
Foreign (non-U.S.) organizations must follow policies described in the NIH Grants Policy Statement , and procedures for foreign organizations described throughout the How to Apply Application Guide.
See Part 2. Section III.1 for information regarding the requirement for obtaining a unique entity identifier and for completing and maintaining active registrations in System for Award Management (SAM), NATO Commercial and Government Entity (NCAGE) Code (if applicable), eRA Commons, and Grants.gov
Part I. contains information about Key Dates and times. Applicants are encouraged to submit applications before the due date to ensure they have time to make any application corrections that might be necessary for successful submission. When a submission date falls on a weekend or Federal holiday , the application deadline is automatically extended to the next business day.
Organizations must submit applications to Grants.gov (the online portal to find and apply for grants across all Federal agencies). Applicants must then complete the submission process by tracking the status of the application in the eRA Commons , NIHs electronic system for grants administration. NIH and Grants.gov systems check the application against many of the application instructions upon submission. Errors must be corrected and a changed/corrected application must be submitted to Grants.gov on or before the application due date and time. If a Changed/Corrected application is submitted after the deadline, the application will be considered late. Applications that miss the due date and time are subjected to the NIH Grants Policy Statement Section 2.3.9.2 Electronically Submitted Applications .
Applicants are responsible for viewing their application before the due date in the eRA Commons to ensure accurate and successful submission.
Information on the submission process and a definition of on-time submission are provided in the How to Apply – Application Guide .
This initiative is not subject to intergovernmental review.
All NIH awards are subject to the terms and conditions, cost principles, and other considerations described in the NIH Grants Policy Statement .
Pre-award costs are allowable only as described in the NIH Grants Policy Statement Section 7.9.1 Selected Items of Cost .
Applications must be submitted electronically following the instructions described in the How to Apply - Application Guide . Paper applications will not be accepted.
Applicants must complete all required registrations before the application due date. Section III. Eligibility Information contains information about registration.
For assistance with your electronic application or for more information on the electronic submission process, visit How to Apply – Application Guide . If you encounter a system issue beyond your control that threatens your ability to complete the submission process on-time, you must follow the Dealing with System Issues guidance. For assistance with application submission, contact the Application Submission Contacts in Section VII.
Important reminders:
All PD(s)/PI(s) must include their eRA Commons ID in the Credential field of the Senior/Key Person Profile form . Failure to register in the Commons and to include a valid PD/PI Commons ID in the credential field will prevent the successful submission of an electronic application to NIH. See Section III of this NOFO for information on registration requirements.
The applicant organization must ensure that the unique entity identifier provided on the application is the same identifier used in the organizations profile in the eRA Commons and for the System for Award Management. Additional information may be found in the How to Apply - Application Guide .
See more tips for avoiding common errors.
Upon receipt, applications will be evaluated for completeness and compliance with application instructions by the Center for Scientific Review and responsiveness by components of participating organizations , NIH. Applications that are incomplete, non-compliant and/or nonresponsive will not be reviewed.
Requests of $500,000 or more for direct costs in any year
Applicants requesting $500,000 or more in direct costs in any year (excluding consortium F&A) must contact a Scientific/ Research Contact at least 6 weeks before submitting the application and follow the Policy on the Acceptance for Review of Unsolicited Applications that Request $500,000 or More in Direct Costs as described in the SF424 (R&R) Application Guide.
Recipients or subrecipients must submit any information related to violations of federal criminal law involving fraud, bribery, or gratuity violations potentially affecting the federal award. See Mandatory Disclosures, 2 CFR 200.113 and NIH Grants Policy Statement Section 4.1.35 .
Send written disclosures to the NIH Chief Grants Management Officer listed on the Notice of Award for the IC that funded the award and to the HHS Office of Inspector Grant Self Disclosure Program at [email protected] .
Applicants are required to follow the instructions for post-submission materials, as described in the policy
1. criteria.
Only the review criteria described below will be considered in the review process. Applications submitted to the NIH in support of the NIH mission are evaluated for scientific and technical merit through the NIH peer review system.
For this particular announcement, note the following:
Immediate clinically translational potential of the proposed project is NOT a requirement for this FOA.
Reviewers will provide an overall impact score to reflect their assessment of the likelihood for the project to exert a sustained, powerful influence on the research field(s) involved, in consideration of the following review criteria and additional review criteria (as applicable for the project proposed).
Reviewers will consider each of the review criteria below in the determination of scientific merit and give a separate score for each. An application does not need to be strong in all categories to be judged likely to have major scientific impact. For example, a project that by its nature is not innovative may be essential to advance a field.
Does the project address an important problem or a critical barrier to progress in the field? Is the prior research that serves as the key support for the proposed project rigorous? If the aims of the project are achieved, how will scientific knowledge, technical capability, and/or clinical practice be improved? How will successful completion of the aims change the concepts, methods, technologies, treatments, services, or preventative interventions that drive this field?
Specific to this NOFO : Does the proposed research project have the potential to advance the understanding of biological mechanisms contributing to cancer health disparities in underrepresented populations?
Are the PD(s)/PI(s), collaborators, and other researchers well suited to the project? If Early Stage Investigators or those in the early stages of independent careers, do they have appropriate experience and training? If established, have they demonstrated an ongoing record of accomplishments that have advanced their field(s)? If the project is collaborative or multi-PD/PI, do the investigators have complementary and integrated expertise; are their leadership approach, governance, and organizational structure appropriate for the project?
Does the application challenge and seek to shift current research or clinical practice paradigms by utilizing novel theoretical concepts, approaches or methodologies, instrumentation, or interventions? Are the concepts, approaches or methodologies, instrumentation, or interventions novel to one field of research or novel in a broad sense? Is a refinement, improvement, or new application of theoretical concepts, approaches or methodologies, instrumentation, or interventions proposed?
Are the overall strategy, methodology, and analyses well-reasoned and appropriate to accomplish the specific aims of the project? Have the investigators included plans to address weaknesses in the rigor of prior research that serves as the key support for the proposed project? Have the investigators presented strategies to ensure a robust and unbiased approach, as appropriate for the work proposed? Are potential problems, alternative strategies, and benchmarks for success presented? If the project is in the early stages of development, will the strategy establish feasibility and will particularly risky aspects be managed? Have the investigators presented adequate plans to address relevant biological variables, such as sex, for studies in vertebrate animals or human subjects?
If the project involves human subjects and/or NIH-defined clinical research, are the plans to address 1) the protection of human subjects from research risks, and 2) inclusion (or exclusion) of individuals on the basis of sex/gender, race, and ethnicity, as well as the inclusion or exclusion of individuals of all ages (including children and older adults), justified in terms of the scientific goals and research strategy proposed?
Specific to this NOFO : Is the scientific approach proposed adequate to effectively study cancer health disparities between at least two populations, in which one or more are underserved?
Will the scientific environment in which the work will be done contribute to the probability of success? Are the institutional support, equipment, and other physical resources available to the investigators adequate for the project proposed? Will the project benefit from unique features of the scientific environment, subject populations, or collaborative arrangements?
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Comparison of cranial cruciate ligament rupture incidence among parous and nulliparous rottweiler bitches: evidence from a lifetime cohort study supporting a paradigm of pregnancy-associated protection against subsequent non-reproductive disease outcomes.
1. introduction, 2. materials and methods, 2.1. study population, 2.2. ascertainment of ccl status, 2.3. reproductive history, reason for nulliparity, 2.4. other exposure variables, 2.4.1. lifetime ovary exposure, 2.4.2. body condition, 2.4.3. dietary pattern, 2.4.4. habitual physical activity, 2.4.5. participation in work/sport activities, 2.4.6. adult height, 2.5. statistical analysis, 3.1. reproductive history and ccl rupture incidence in the overall study sample, 3.2. comparison of characteristics of bitches in the nulliparous and parous groups, 3.3. parity, the production of live offspring, is associated with significant ccl rupture risk reduction, 3.4. the main finding, the association between parity and ccl rupture risk reduction, is not explained by owner decision on why nulliparous females did not become parous, 4. discussion, 5. conclusions, author contributions, institutional review board statement, informed consent statement, data availability statement, acknowledgments, conflicts of interest, abbreviations.
Click here to enlarge figure
Variable | Females from the Exceptional Aging in Rottweilers Study | |||
---|---|---|---|---|
Total (n = 65) | Nulliparous (n = 33) | Parous (n = 32) | p-Value | |
# of U.S. states, n | 30 states and Canada | 17 states and Canada | 20 states and Canada | -- |
# of households, n | 61 | 33 | 31 | -- |
1985–1997, n (%) | 23 (35.4) | 10 (30.3) | 13 (40.6) | 0.38 |
1998–2010, n (%) | 42 (64.6) | 23 (69.7) | 19 (59.4) | |
(years), median (IQR) | 14.0 (13.6, 14.5) | 14.0 (13.7, 14.4) | 13.9 (13.5, 14.7) | 0.79 |
(years), median (IQR) | 5.8 (3.9, 7.1) | 4.3 (3.2, 5.9) | 6.8 (5.6, 8.0) | <0.001 |
(cm), median (IQR) | 58.4 (57.2, 61.0) | 58.4 (57.2, 60.8) | 61.0 (58.4, 62.2) | 0.07 |
, n (%) | 22 (33.8) | 10 (30.3) | 12 (37.5) | 0.54 |
Never, n (%) | 35 (53.8) | 21 (63.6) | 14 (43.8) | 0.11 |
Ever, n (%) | 30 (46.2) | 12 (36.4) | 18 (56.3) | |
Low risk, n (%) | 44 (67.7) | 20 (60.6) | 24 (75.0) | 0.22 |
High risk, n (%) | 21 (32.3) | 13 (39.4) | 8 (25.0) | |
No work/sport activity, n (%) | 37 (56.9) | 18 (54.5) | 19 (59.4) | 0.69 |
Work/sport activity, n (%) | 28 (43.1) | 15 (45.5) | 13 (40.6) | |
Any CCL rupture, n (%) | 17 (26.2) | 14 (42.4) | 3 (9.4) | 0.004 |
Bilateral CCL rupture, n (% of CCL ruptures) | 10 (58.8) | 9 (64.3) | 1 (33.3) | 0.005 |
Age at first rupture (years), median (IQR) | 6.5 (4.9, 8.2) | 5.8 (4.4, 8.4) | 7.9 (7.7, n/a) | 0.19 |
CCL rupture incidence rate (95% CI) expressed as cases per 10,000 DYAR | 237 (138–380) | 427 (233–716) | 77 (16–226) | 0.003 |
Variable | Unadjusted OR (95% CI) | p-Value | Adjusted OR (95% CI) | p-Value |
---|---|---|---|---|
Non-parous | 1.0 (ref) | -- | -- | -- |
Parous | 0.14 (0.04–0.56) | 0.005 | 0.06 (0.01–0.46) | 0.006 |
≥5.8 years | 1.0 (ref) | -- | -- | -- |
<5.8 years | 2.36 (0.75–7.42) | 0.14 | 0.97 (0.21–4.60) | 0.97 |
Precedent Overweight | 1.0 (ref) | -- | -- | -- |
Not overweight | 1.34 (0.37–4.82) | 0.66 | 0.52 (0.11–2.47) | 0.41 |
Never | 1.0 (ref) | -- | -- | -- |
Ever | 1.45 (0.48–4.39) | 0.52 | 2.38 (0.60–9.47) | 0.22 |
Low risk | 1.0 (ref) | -- | -- | -- |
High risk | 0.83 (0.25–2.78) | 0.77 | 0.34 (0.07–1.58) | 0.17 |
No work/sport activity | 1.0 (ref) | -- | -- | -- |
Work/sport activity | 3.34 (1.05–10.64) | 0.04 | 5.43 (1.27–23.51) | 0.02 |
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Carrillo, A.E.; Maras, A.H.; Suckow, C.L.; Chiang, E.C.; Waters, D.J. Comparison of Cranial Cruciate Ligament Rupture Incidence among Parous and Nulliparous Rottweiler Bitches: Evidence from a Lifetime Cohort Study Supporting a Paradigm of Pregnancy-Associated Protection against Subsequent Non-Reproductive Disease Outcomes. Animals 2024 , 14 , 2608. https://doi.org/10.3390/ani14172608
Carrillo AE, Maras AH, Suckow CL, Chiang EC, Waters DJ. Comparison of Cranial Cruciate Ligament Rupture Incidence among Parous and Nulliparous Rottweiler Bitches: Evidence from a Lifetime Cohort Study Supporting a Paradigm of Pregnancy-Associated Protection against Subsequent Non-Reproductive Disease Outcomes. Animals . 2024; 14(17):2608. https://doi.org/10.3390/ani14172608
Carrillo, Andres E., Aimee H. Maras, Cheri L. Suckow, Emily C. Chiang, and David J. Waters. 2024. "Comparison of Cranial Cruciate Ligament Rupture Incidence among Parous and Nulliparous Rottweiler Bitches: Evidence from a Lifetime Cohort Study Supporting a Paradigm of Pregnancy-Associated Protection against Subsequent Non-Reproductive Disease Outcomes" Animals 14, no. 17: 2608. https://doi.org/10.3390/ani14172608
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Jianhui xiao.
1 Department of Geriatrics, Shenzhen Hospital of Integrated Traditional Chinese and Western Medicine, Shenzhen, China
2 Department of Emergency Medicine, the First Affiliated Hospital of Shenzhen University, Shenzhen Second People’s Hospital, Shenzhen, China
Xiang Hu, First Affiliated Hospital of Wenzhou Medical University, China
Publicly available datasets were analyzed in this study. This data can be found here: https://bmjopen.bmj.com/content/8/9/e021768 .
Given the limited evidence on the relationship between the triglyceride-glucose (TyG) index and the risk of prediabetes among young adults, our study aimed to investigate the potential impact of the TyG index on the future development of prediabetes in young individuals.
This retrospective cohort study included 125,327 healthy adults aged 20 to 45 years. We utilized Cox proportional hazards regression models, combined with cubic spline functions and smooth curve fitting, to assess the relationship between baseline TyG index and the risk of prediabetes among young adults, exploring its non-linear association. A series of sensitivity analyses and subgroup analyses were conducted to ensure the robustness of our findings.
After adjusting for covariates, the study found a positive correlation between the TyG index and the risk of prediabetes (HR=1.81, 95%CI: 1.54–2.13, p<0.0001). The risk of prediabetes increased progressively across quartiles of the TyG index (Q1 to Q4), with Q4 showing a significantly higher risk compared to Q1 (adjusted HR=2.33, 95% CI=1.72–3.16). Moreover, a non-linear relationship was identified between the TyG index and the risk of prediabetes, with an inflection point at 9.39. To the left of the inflection point, the HR was 2.04 (95% CI: 1.69 to 2.46), while to the right, the HR was 0.89 (95% CI: 0.48 to 1.65).
Our study reveals a non-linear relationship and a saturation effect between the TyG index and the development of prediabetes among young individuals in China, with an inflection point at 9.39. Understanding this non-linear relationship can assist clinicians in identifying young individuals at high risk and implementing targeted interventions to reduce their risk of progressing to diabetes.
Prediabetes is characterized by elevated blood glucose levels below the diagnostic threshold for diabetes but is associated with a higher risk of developing diabetes ( 1 ). The standardized prevalence of prediabetes in Chinese adults was 35.2% (33.5% to 37.0%) ( 2 ). Approximately 70% of prediabetes patients will progress to diabetes within 10 years, and the incidence of diabetes exceeds 90% within 20 years ( 3 ). Prediabetes is also associated with a high burden of cardiometabolic risk factors and poor outcomes ( 4 ). In a large meta-analysis of prospective studies (53 studies, 1.6 million individuals, median follow-up duration 9.5 years) examining the risks of cardiovascular disease and death in persons with prediabetes compared with normal glycemia ( 5 ), prediabetes was associated with an increased risk of cardiovascular disease and all-cause mortality. Several cohort studies have shown that individuals with prediabetes have a higher all-cause mortality rate compared to those with normal glycemia ( 6 – 12 ). Similarly, individuals with prediabetes have a high risk of hospitalization ( 11 ). Age is a significant risk factor for prediabetes, with evidence showing a strong association between the increase in prediabetes and age. In a 2011–2012 National Health and Nutrition Examination Survey (NHANES) analysis, in adults, the prevalence of prediabetes was 38.0%, while among young individuals aged 20–44 years, it reached up to 28.2% ( 13 ). Prediabetes is commonly considered a warning sign; nevertheless, many prediabetic patients, particularly younger individuals, tend to neglect this metabolic abnormality and underestimate its importance.
The triglyceride-glucose (TyG) index is calculated as the product of fasting plasma glucose (FPG) levels and triglyceride (TG) ( 14 ). Due to its convenience and ease of calculation, the TyG index has been widely used in various clinical settings ( 15 – 20 ). Insulin resistance (IR) is a key factor in metabolic disorders such as metabolic syndrome, non-alcoholic fatty liver disease (NAFLD), diabetes, and obesity ( 21 , 22 ). IR is a major contributor to the progression from normal glucose resistance to prediabetes and then to diabetes. The TyG index is considered a novel marker of IR that accurately and reliably reflects the extent of IR ( 23 , 24 ). Evidence suggests an association between the TyG index and the prevalence of prediabetes in middle-aged and older adults, but it is unclear whether this association exists in young adults ( 25 ).
Due to economic growth, lifestyle, and diet changes, diabetes rates are rising, with younger people increasingly affected. Young diabetics may show unusual symptoms and be overlooked, making early risk identification and intervention critical ( 26 , 27 ). This study used existing data to examine the TyG index and prediabetes risk in individuals aged 20–45, aiming to support its clinical use for early prediabetes detection.
This study utilized data from a previous retrospective cohort study conducted by Chinese researchers (Chen et al.) ( 28 ). The target independent variable was the TyG at baseline. The outcome variable was the development from normoglycemia to prediabetes at follow-up.
Access to the original dataset was granted at no cost through the DATADRYAD platform ( www.datadryad.org ), courtesy of Ying Chen et al. In accordance with Dryad’s usage policy, the data is available for academic and research purposes, allowing users to share, adapt, alter, and build upon the material, provided it is not for commercial use and proper attribution is given to the original authors and source. The dataset was sourced from a publicly accessible study published in 2018 titled “Association of body mass index and age with diabetes onset in Chinese adults: a population-based cohort study,” which can be found at http://dx.doi.org/10.1136/bmjopen-2018–021768 . For those interested, the dataset can be retrieved from the following link: https://doi.org/10.5061/dryad.ft8750v ( 28 ). Given that the current study involves a secondary analysis of existing data, there was no need for obtaining informed consent or additional ethical approval.
The primary research included 685,277 healthy individuals aged 20 years and above, all of whom had undergone a minimum of two health assessments. The study focused on participants who, during follow-up, had FPG levels ranging from 110 to 125 mg/dL without any prior diagnosis of diabetes. The initial selection excluded participants based on several factors (1): lack of detailed information regarding weight, height, or gender (2); BMI values outside the normal range (<15 kg/m 2 or >55 kg/m 2 ) (3); intervals between visits shorter than 2 years (4); missing fasting plasma glucose readings (5); individuals diagnosed with diabetes at the start or with uncertain diabetes status at the time of follow-up. Following these criteria, the study retained 211,833 participants.
Further analysis led to the exclusion of an additional 86,506 participants for reasons including: 1) absence of follow-up FPG readings, 2) baseline FPG levels ≥100 mg/dL, 3) diabetes diagnosis at follow-up, 4) had no available TG value, and 5) lack of TG values or being over 45 years of age. Young people are defined as those aged 45 years old or younger ( 29 ). Ultimately, the study included 125,327 healthy participants. The process of selecting participants for this study is depicted in Figure 1 .
Flowchart illustrating the selection process of study participants.
In this study, data collection included demographic information such as age, systolic blood pressure (SBP), diastolic blood pressure (DBP), height, and weight, from which body mass index (BMI) was calculated. To ensure consistency in data collection, staff received specialized training focusing on demographic data and key measurements, including blood pressure. Tests were uniformly conducted in a standardized laboratory environment for FPG, serum creatinine (Scr), TG, total cholesterol (TC), blood urea nitrogen (BUN), alanine aminotransferase (ALT), low-density lipoprotein cholesterol (LDL-c), and high-density lipoprotein cholesterol (HDL-c). Additionally, the study collected information on the patients’ smoking and drinking histories, defining current drinking as 1, former drinking as 2, never drinking as 3. Similarly, current smoking was coded as 1, former smoking as 2, never smoking as 3.
Our primary outcome was the occurrence of prediabetes, defined by FPG levels in the range of 100–125 mg/dL at follow-up without reported incident diabetes ( 30 ).
The TyG index, the primary exposure variable in this study, is defined as follows: TyG index = ln [FPG (mg/dL) × TG (mg/dL)/2] ( 31 ). We divided it into four quartiles and considered it as a continuous variable. For continuous variables following a normal distribution, we reported the mean and standard deviation; for non-normally distributed data, we provided the median. For categorical variables, we presented the frequency and proportion. To compare differences between TyG index groups, we used the Kruskal-Wallis H test (for skewed distributed data), one-way analysis of variance (for normally distributed data), or chi-square test (for categorical variables).
We constructed several models to assess the relationship between the TyG index and prediabetes risk: a baseline model without any adjustments, a simplified model adjusting for gender and age only (Model I), and a comprehensive model adjusting for multiple covariates (Model II: including age, gender, BMI, systolic blood pressure, diastolic blood pressure, alanine aminotransferase, urea nitrogen, aspartate aminotransferase, total cholesterol, high-density lipoprotein cholesterol, low-density lipoprotein cholesterol, serum creatinine, family history of diabetes, alcohol consumption, and smoking status). We recorded the effect size (hazard ratio, HR) and its 95% confidence interval (CI) for each model.
We adjusted for potential confounding factors based on clinical experience, literature review, and univariate analysis results. Additionally, we used a multivariable Cox proportional hazards model and introduced cubic spline functions and smooth curve fitting to explore the possible nonlinear relationship between the TyG index and prediabetes risk. We also used a segmented Cox proportional hazards model to further clarify this nonlinear relationship. Unmeasured confounding between TyG and Prediabetes risk was assessed by calculating E-values.
To validate our findings, we conducted a series of sensitivity analyses. By incorporating continuous variables into a generalized additive model (GAM) in curve form, we further confirmed the robustness of the results. Additionally, we conducted analyses using stratified Cox proportional hazards models in different subgroups (such as age, gender, blood pressure, smoking, and drinking status). Finally, we used likelihood ratio tests to examine whether there were interactions in the model, both in models including interaction terms and those without. All analyses were performed using Empower Stats (X&Y Solutions, Inc., Boston, MA, http://www.empowerstats.com ), with a statistical significance level set at a two-sided P value less than 0.05.
We conducted a comprehensive analysis of the baseline characteristics of participants to investigate the relationship between the TyG index and the risk of prediabetes. The results are summarized in Table 1 , demonstrating significant differences between normal and prediabetes groups across various parameters. The median follow-up duration of our study was 3.00 years, ranging from 2.00 to 6.20 years. The prediabetes group exhibited a slightly higher mean age (35.70 years) compared to the normal group (34.31 years), along with increased height, weight, and BMI. SBP and DBP were also notably elevated in the prediabetes group. The prediabetes group displayed higher levels of FPG, TG, and the TyG index, indicating a potential link between these parameters and the development of prediabetes. Additionally, AST, ALT, BUN, and Scr levels were higher in the prediabetes group. In addition, higher percentage of males, current smokers, and current drinkers were observed in the prediabetes group, along with a greater prevalence of family history of diabetes. Furthermore, the average follow-up duration was slightly longer in the prediabetes group compared to the normal group. As illustrated in Figure 2 , a detailed Distribution of TyG index was conducted. It presented a normal distribution, ranging from 4.65 to 11.78, with a mean of 8.23.
The baseline characteristics of participants.
TyG index | Normal | prediabetes | P-value |
---|---|---|---|
Participants | 114,999 | 10,328 | |
Age (years) | 34.31 ± 5.31 | 35.70 ± 5.32 | <0.001 |
Height (cm) | 166.98 ± 8.30 | 168.75 ± 8.13 | <0.001 |
Weight (kg) | 63.15 ± 12.39 | 68.90 ± 13.03 | <0.001 |
BMI (kg/m ) | 22.52 ± 3.27 | 24.08 ± 3.57 | <0.001 |
SBP (mmHg) | 114.74 ± 13.85 | 120.30 ± 14.59 | <0.001 |
DBP (mmHg) | 71.66 ± 9.77 | 75.11 ± 10.50 | <0.001 |
FPG at baseline (mg/dL) | 84.96 ± 8.59 | 90.10 ± 7.65 | <0.001 |
TG (mg/dL) | 102.15 ± 73.75 | 130.38 ± 98.60 | <0.001 |
TyG index | 8.21 ± 0.56 | 8.48 ± 0.61 | <0.001 |
ALT (U/L) | 16.50 (12.00–26.00) | 22.00 (14.10–35.80) | <0.001 |
AST (U/L) | 21.00 (18.00–25.20) | 22.30 (19.00–28.00) | <0.001 |
BUN (mmol/L) | 4.46 ± 1.11 | 4.65 ± 1.12 | <0.001 |
Scr (μmol/L) | 68.90 ± 15.10 | 72.65 ± 14.81 | <0.001 |
TC (mmol/L) | 4.52 ± 0.83 | 4.65 ± 0.85 | <0.001 |
HDL-c (mmol/L) | 1.38 ± 0.30 | 1.32 ± 0.29 | <0.001 |
LDL-c (mmol/L) | 2.63 ± 0.63 | 2.72 ± 0.63 | <0.001 |
Sex | <0.001 | ||
Male | 59,197 (51.48%) | 7,039 (68.15%) | |
Female | 55,802 (48.52%) | 3,289 (31.85%) | |
Smoking status | <0.001 | ||
Current smoker | 4,684 (14.19%) | 620 (19.40%) | |
Ever smoker | 1,448 (4.39%) | 169 (5.29%) | |
Never | 26,882 (81.43%) | 2,407 (75.31%) | |
Drinking status | <0.001 | ||
Current drinker | 410 (1.24%) | 60 (1.88%) | |
Ever drinker | 4,801 (14.54%) | 602 (18.84%) | |
Never | 27,803 (84.22%) | 2,534 (79.29%) | |
Family history of diabetes | <0.001 | ||
No | 112,693 (97.99%) | 10,031 (97.12%) | |
Yes | 2306 (2.01%) | 297 (2.88%) | |
Follow-up (year) | 3.12 ± 0.93 | 3.23 ± 0.96 | <0.001 |
Continuous variables were summarized as mean (SD) or medians (quartile interval); categorical variables were displayed as percentage (%).
BMI, body mass index; SBP, systolic blood pressure; DBP; diastolic blood pressure; TG triglyceride; TC, total cholesterol; HDL-c, high-density lipoprotein cholesterol; LDL-c, low-density lipoprotein cholesterol; AST aspartate aminotransferase; ALT, alanine aminotransferase; BUN, blood urea nitrogen; Scr, serum creatinine; FPG, fasting plasma glucose; TyG index, triglyceride glucose index.
Distribution of TyG index. It presented a normal distribution, ranging from4.65 to 11.78, with a mean of 8.23.
Table 2 and Figure 3 describe the incidence rates of prediabetes. Among the participants, 10,328 (8.24%) developed prediabetes. Participants were divided into subgroups based on the quartiles of the TyG index. The incidence rates of prediabetes per 1000 person-years were 14.426, 20.012, 27.066, and 44.490 for each TyG index quartile. The incidence rates of prediabetes in each TyG index quartile were as follows: Q1: 4.68%, Q2: 6.27%, Q3: 8.33%, and Q4: 13.69%. Participants with the highest TyG index (Q4) had a higher risk of developing prediabetes compared to those with the lowest TyG index (Q1) (trend p < 0.001).
The Incidence rate of prediabetes (% or Per 1,000 person-year).
TyG index (quartile) | Participants (n) | Prediabetes events (n) | Incidence rate (95%CI) (%) | Per 1,000 person-year |
---|---|---|---|---|
Total | 125327 | 10328 | 8.24 (8.09–8.39) | 26.355 |
Q1 | 31,328 | 1466 | 4.68 (4.45–4.91) | 14.426 |
Q2 | 31,317 | 1962 | 6.27 (6.00–6.53) | 20.012 |
Q3 | 31,349 | 2611 | 8.33 (8.02–8.64) | 27.066 |
Q4 | 31,332 | 4289 | 13.69 (13.31–14.07) | 44.490 |
P for trend | <0.001 |
Kaplan–Meier event-free survival curve. Kaplan–Meier analysis of incident prediabetes based on two group (log-rank, P < 0.0001).
In Table 3 , the relationship between the TyG index and the risk of prediabetes is presented across different models. The results are as follows: In the crude model, the HR for the TyG index was 2.17 (95%CI 2.11–2.24, P < 0.0001). After adjusting for age and sex (Model I), the HR decreased to 1.94 (95%CI 1.88–2.00, P < 0.0001). Further adjustments in Model II, which included age, sex, SBP, DBP, BMI, ALT, AST, BUN, Scr, TC, LDL-C, HDL-c, family history of diabetes, drinking status, and smoking status, resulted in an HR of 1.81 (95%CI 1.54–2.13, P < 0.0001).
Relationship between TyG index and risk of prediabetes in different models.
Exposure | Crude model (HR,95%CI) P | Model I (HR,95%CI) P | Model II (HR,95%CI) P | Model III (HR,95%CI) P |
---|---|---|---|---|
TyG index | 2.17 (2.11, 2.24) <0.0001 | 1.94 (1.88, 2.00) <0.0001 | 1.81 (1.54, 2.13) <0.0001 | 1.86 (1.57, 2.19) <0.0001 |
(TyG index quartiles) | ||||
Q1 | Ref | Ref | Ref | Ref |
Q2 | 1.49 (1.39, 1.59) <0.0001 | 1.40 (1.31, 1.50) <0.0001 | 1.29 (0.95, 1.75) 0.1065 | 1.30 (0.95, 1.77) 0.0998 |
Q3 | 2.11 (1.98, 2.25) <0.0001 | 1.86 (1.74, 1.99) <0.0001 | 1.75 (1.30, 2.34) 0.0002 | 1.81 (1.35, 2.44) <0.0001 |
Q4 | 3.56 (3.36, 3.78) <0.0001 | 2.90 (2.72, 3.09) <0.0001 | 2.33 (1.72, 3.16) <0.0001 | 2.44 (1.78, 3.34) <0.0001 |
P for trend | <0.0001 | <0.0001 | <0.0001 | <0.0001 |
Crude model: we did not adjust other covariates.
Model I: we adjusted age, sex.
Model II: we adjusted age, sex, SBP, DBP, BMI, ALT, AST, BUN, Scr, TC, LDL-C, HDL-c, family history of diabetes, drinking status, and smoking status.
Model III: we adjusted age(smooth), sex, SBP (smooth), DBP (smooth), BMI (smooth), BUN (smooth), Scr (smooth), ALT (smooth), AST (smooth), TC (smooth), LDL-C(smooth), HDL-c(smooth), smoking status, drinking status, family history of diabetes.
HR, Hazard ratios; CI, confidence, Ref, reference.
After adjustments, similar trends were observed across all quartiles, indicating a significant association between the TyG index quartiles and the risk of prediabetes (P for trend < 0.0001 for all models).
We performed a series of sensitivity analyses to confirm the reliability of our conclusions. Initially, the generalized additive model (GAM) in Model III, which included additional smooth terms for various variables, showed an HR of 1.86 (95%CI 1.57–2.19, P < 0.0001) ( Table 3 ). Secondly, participants with a BMI ≥ 28 kg/m 2 were excluded ( 32 ). After controlling for confounding variables, the results demonstrated a persistent positive association between the TyG index and the risk of prediabetes (HR= 1.82, 95% CI: 1.52–2.17, p < 0.0001). Furthermore, we performed a sensitivity analysis excluding patients aged ≥ 40 years. After adjusting for confounding variables, the results continued to demonstrate a positive association between the TyG index and the risk of prediabetes (HR = 1.99, 95% CI: 1.64–2.42, p < 0.0001). Additionally, we conducted an analysis on participants without a family history of diabetes, revealing a risk ratio of 1.83 (95% confidence interval: 1.54–2.16, p < 0.0001). Based on all sensitivity analyses, we concluded that our findings are reliable ( Table 4 ).
Relationship between TyG index and the risk of prediabetes in different sensitivity analyses.
Exposure | Model I (HR,95%CI) P | Model II (HR,95%CI) P | Model III (HR,95%CI) P |
---|---|---|---|
TyG index | 1.82 (1.52, 2.17) <0.0001 | 1.99 (1.64, 2.42) <0.0001 | 1.83 (1.54, 2.16) <0.0001 |
(TyG index quartiles) | |||
Q1 | Ref | Ref | Ref |
Q2 | 1.33 (0.97, 1.82) 0.0792 | 1.18 (0.84, 1.65) 0.3502 | 1.21 (0.88, 1.67) 0.2386 |
Q3 | 1.77 (1.31, 2.40) 0.0002 | 1.59 (1.15, 2.20) 0.0053 | 1.67 (1.23, 2.26) 0.0010 |
Q4 | 2.39 (1.73, 3.29) <0.0001 | 2.14 (1.52, 3.01) <0.0001 | 2.25 (1.64, 3.09) <0.0001 |
P for trend | <0.0001 | <0.0001 | <0.0001 |
Model I was a sensitivity analysis performed after excluding participants with BMI≥ 28 kg/m 2 (N= 8,595). we adjusted age, sex, BMI, SBP, DBP, ALT, AST, BUN, Scr, TC, LDL-C, HDL-c, family history of diabetes, drinking status, and smoking status.
Model II was a sensitivity analysis performed after excluding participants with age≥ 40 years (N= 25,255). we adjusted age, sex, BMI, SBP, DBP, ALT, AST, BUN, Scr, TC, LDL-C, HDL-c, family history of diabetes, drinking status, and smoking status.
Model III was a sensitivity analysis performed on participants without family of diabetes (N= 2,603). We adjusted age, sex, SBP, DBP, BMI, ALT, AST, BUN, Scr, TC, LDL-C, HDL-c, drinking status, and smoking status. HR, Hazard ratios; CI, confidence, Ref, reference.
We identified a non-linear relationship between the TyG index and the risk of prediabetes ( Figure 4 and Table 5 ). Firstly, we used a Cox proportional hazards regression model with cubic spline functions to evaluate the relationship between the TyG index and prediabetes risk. The result showed that the relationship between the TyG index and prediabetes risk was non-linear. We employed a two-piecewise Cox proportional hazards regression model to investigate the relationship between the TyG index and the risk of developing prediabetes. The standard Cox regression model revealed a HR of 1.81 (95% CI: 1.54, 2.13) with a P value of <0.0001, indicating a significant association between the TyG index and prediabetes risk. In addition, we identified an inflection point in the TyG index at 9.39. Below this inflection point (<9.39), there is a significant positive correlation between the TyG index and the risk of prediabetes (HR: 2.04, 95% CI: 1.69, 2.46, P <0.0001). Conversely, Conversely, when the TyG index is 9.39 or higher, their relationship is not statistically significant (HR: 0.89, 95% CI: 0.48, 1.65, P=0.7019).
The non-linear relationship between TyG index and the risk of prediabetes in participants. We used a Cox proportional hazards regression model with cubic spline functions to evaluate the relationship between the TyG index and prediabetes risk. We adjusted age, sex, SBP, DBP, BMI, ALT, AST, BUN, Scr, TC, LDL-C, HDL-c, family history of diabetes, drinking status, and smoking status. The result showed that the relationship between the TyG index and prediabetes risk was non-linear, with the inflection point of TyG index being 9.39.
The result of the two-piecewise Cox proportional hazards regression model.
Outcome: prediabetes | HR, 95% CI | P-value |
---|---|---|
Fitting model by standard Cox regression | 1.81 (1.54, 2.13) | <0.0001 |
Fitting model by two-piecewise Cox regression | ||
Inflection points of TyG index | 9.39 | |
<9.39 | 2.04 (1.69, 2.46) | <0.0001 |
≥9.39 | 0.89 (0.48, 1.65) | 0.7019 |
P for log likelihood ratio test | 0.011 |
We adjusted age, sex, SBP, DBP, BMI, ALT, AST, BUN, Scr, TC, LDL-C, HDL-c, family history of diabetes, drinking status, and smoking status.
As illustrated in Table 6 , a detailed subgroup analysis was conducted. Gender, age, BMI, systolic and diastolic blood pressures, smoking and drinking habits, and a family history of diabetes did not alter the association between the TyG index and the risk of prediabetes. Thus, no significant interactions were observed between these variables and the TyG index (all interaction P > 0.05).
Effect size of TyG index on prediabetes in prespecified and exploratory subgroups.
Variable | HR (95% CI) | P-value | P for interaction |
---|---|---|---|
0.4260 | |||
20–30 | 2.14 (1.40, 3.25) | 0.0004 | |
30–40 | 1.90 (1.57, 2.30) | <0.0001 | |
>40 | 1.66 (1.32, 2.08) | <0.0001 | |
0.7626 | |||
<18 | 2.73 (0.55, 13.53) | 0.2186 | |
18–24 | 1.95 (1.56, 2.42) | <0.0001 | |
24–28 | 1.71 (1.38, 2.14) | <0.0001 | |
≥28 | 1.95 (1.39, 2.76) | 0.0001 | |
0.1758 | |||
Male | 1.74 (1.46, 2.06) | <0.0001 | |
Female | 2.19 (1.59, 3.03) | <0.0001 | |
0.660 | |||
<140 | 1.79 (1.52, 2.12) | <0.0001 | |
≥140 | 2.00 (1.25, 3.20) | 0.0040 | |
0.3484 | |||
<90 | 1.79 (1.52, 2.11) | <0.0001 | |
≥90 | 2.22 (1.42, 3.47) | 0.0005 | |
0.3930 | |||
Current drinker | 2.93 (1.20, 7.14) | 0.0183 | |
Ever drinker | 1.66 (1.30, 2.12) | <0.0001 | |
Never | 1.87 (1.55, 2.24) | <0.0001 | |
0.2937 | |||
Current smoker | 1.58 (1.22, 2.06) | 0.0006 | |
Ever smoker | 2.38 (1.43, 3.96) | 0.0008 | |
Never | 1.86 (1.55, 2.23) | <0.0001 | |
0.2497 | |||
Yes | 1.86 (1.57, 2.19) | <0.0001 | |
No | 1.49 (1.03, 2.16) | 0.0360 |
Above model adjusted for age, sex, SBP, DBP, BMI, ALT, AST, BUN, Scr, TC, LDL-C, HDL-c, family history of diabetes, drinking status, and smoking status. In each case, the model is not adjusted for the stratification variable when the stratification variable was a categorical variable.
This retrospective cohort study revealed that a higher TyG index independently predicts an increased risk of prediabetes in Chinese adults aged 20–45. Furthermore, a nonlinear relationship was observed, indicating that the risk of prediabetes tends to rise with an increasing TyG index. Individuals in the highest TyG index quartile had a 2.33-fold increased risk of developing prediabetes compared to those in the lowest quartile. Additionally, a threshold effect curve was identified, revealing varying relationships between the TyG index and prediabetes risk across the inflection point.
Initially, the TyG index has been widely validated as an indicator of IR in various epidemiological studies. A cross-sectional study conducted in Mexico involving 99 individuals identified the TyG index as an optimal tool for assessing IR, showing a high sensitivity (96.5%) and good specificity (85.0%) compared to the gold standard, HIEC ( 23 ). Research involving 82 Brazilian subjects corroborated the TyG index as a more precise predictor of IR compared to HOMA-IR ( 33 ). Numerous studies have indicated an association between the TyG index and cardiovascular disease (CVD) risk ( 34 – 37 ). A study involving 4,340 American patients under the age of 65 with prediabetes or diabetes found that an increase in the TyG index is associated with a higher incidence of CVD, including congestive heart failure (CHF), coronary heart disease (CHD), atherosclerotic cardiovascular disease (ASCVD), heart attack, angina, and stroke ( 34 ). Additionally, several studies have found a close relationship between the TyG index and coronary artery disease ( 35 – 37 ). Furthermore, recent studies have identified a significant correlation between the TyG index and both cognitive and physical impairments in elderly individuals with prefrail hypertension ( 38 ).
Recently, there has been a growing body of research exploring the relationship between the TyG index and the risk of diabetes ( 39 – 42 ). A prospective cohort study conducted in rural China suggests that an increasing TyG index is associated with a higher cumulative risk of developing incident T2DM among individuals with normal weight ( 43 ). Another retrospective cohort study demonstrated a link between elevated TyG index levels and an augmented risk of diabetes within the Chinese population ( 44 ). Additionally, recent studies have indicated that there is a positive correlation between the TyG index and the risk of diabetes ( 45 ). It is noteworthy that prediabetes is a critical period for promoting, preventing, or delaying the development of diabetes mellitus (DM).
But research on the TyG index’s association with prediabetes is limited. An initial study demonstrated that the TyG index is comparable to HbA1C as a diagnostic marker for prediabetes ( 46 ). Findings from a case-control study conducted in Palembang, Indonesia, involving 570 participants without a family history of T2DM (265 prediabetes cases and an equal number of age-matched controls), indicated that the TyG index was the primary risk predictor for prediabetes ( 47 ). A prospective cohort study in China indicated that the predictive capacity of the TyG index in forecasting prediabetes surpassed that of obesity, lipid profiles, and other non-insulin-based IR indices ( 48 ). A cross-sectional study conducted in the United States, involving 25,159 participants, found that 23.88% had prediabetes and 16.22% were diagnosed with diabetes. The study revealed a positive relationship between the TyG index and the prevalence of prediabetes and diabetes ( 49 ). Interestingly, a cohort study conducted in China indicated that, among prediabetic patients, an elevated TyG-BMI might reduce the likelihood of returning to normal glucose levels in the future ( 50 ). However, no specific studies have investigated the association between TyG index and prediabetes in a young population.
Therefore, we hypothesize that the TyG index is positively correlated with the risk of prediabetes in young populations. Therefore, our study investigates the TyG index’s role in identifying prediabetes risk factors among young individuals, aiming to enhance prevention strategies for diabetes onset and progression within this demographic. Initially, our findings indicate the TyG index as a reliable predictor of prediabetes and an independent risk factor, consistent with existing research [45–46]. Furthermore, our study has shown a positive correlation between the TyG index and prediabetes, consistent with our research hypothesis. That is for every 1-unit increase in the TyG index, the risk of prediabetes will increase by 81% (HR: 1.81, 95% CI 1.54–2.13, P < 0.0001). Similarly, a cohort study conducted in China also identified a positive correlation between TyG and the risk of prediabetes after adjusting for variables ( 48 ). However, there are significant differences between the two studies in terms of subjects, outcomes, and adjusted variables. Their study sample is relatively small compared to ours. They also adjusted for different variables, such as age, cigarette smoking, alcohol consumption, education level, family history of diabetes, hypertension, and cardiovascular disease history, which differs slightly from our study. Additionally, they did not use the Cox proportional hazards regression model combined with cubic splines and smoothing curve fits to examine the nonlinear association between the TyG index and the risk of prediabetes. A study in the United States also found a nonlinear relationship between the TyG index and the prevalence of prediabetes and diabetes, identifying an inflection point at 8 ( 49 ), which is very similar to our conclusion but with some notable differences. Firstly, their study was cross-sectional, which has limitations such as a smaller sample size, unclear causality, and selection bias. Secondly, there are differences in the age distribution of the study populations between the two studies. Additionally, variations in the adjusted variables between the two studies may impact the research outcomes. Although both studies confirm a nonlinear relationship, we identified a critical threshold of 9.39. Our study focused on prediabetes as the outcome, while their study included both prediabetes and diabetes, which could be the primary reason for the earlier inflection point in their findings.
Our study has significant clinical implications. We have identified a non-linear relationship with a saturation effect between the TyG index and the risk of prediabetes in young individuals. Monitoring TyG index values can facilitate early interventions in young individuals to decrease the occurrence of prediabetes. Specifically, when the TyG index is below 9.39, by lowering FBG and TG, we can significantly reduce the risk of prediabetes. However, when the TyG index exceeds 9.39, merely controlling FBG and TG is insufficient to reduce the risk of prediabetes. Thus, it is necessary to manage other related risk factors such as BMI, smoking, and alcohol consumption to better prevent the onset of the disease. Consequently, it is advisable for young people to start making lifestyle changes early, including reducing the intake of high-fat foods, increasing physical activity, and controlling blood sugar levels. These measures can effectively help reduce the risk of developing prediabetes.
This study has some advantages. Firstly, we first identified a nonlinear relationship between the TyG index and prediabetes risk in young individuals. Furthermore, this study included a large sample of 125,327 young adults, minimizing potential biases. Sensitivity analysis was conducted to ensure the reliability and robustness of the results, and a two-stage Cox proportional hazards regression model was applied to determine the inflection point of the TyG index. These findings underscore the critical threshold between the TyG index and diabetes risk, emphasizing the importance of TyG index monitoring and early intervention strategies in diabetes prevention efforts. Additionally, subgroup analysis and interaction tests were performed. The results revealed variations in the impact of the TyG index on prediabetes risk across different subgroups, yet interaction analysis did not indicate significant interactions between the TyG index and factors such as drinking status, smoking status, or family history of diabetes. This further validates the stability of the results.
This study has several limitations. Firstly, as all participants were of Chinese descent, further research is needed to determine the relationship between the TyG index and the risk of prediabetes in individuals with different genetic backgrounds. Secondly, since this large-scale cohort study only focused on young adults in China, these findings may not be applicable to other racial groups and specific populations, such as children and the elderly. Thirdly, the average follow-up time for participants was relatively short. Future research could benefit from extending follow-up duration to mitigate potential misinterpretations or chance findings from short-term observations. Fourthly, there was no data available on serum insulin levels to compare the predictive value of the TyG index and HOMA-IR; additionally, there was also no data on the use of lipid-lowering medications, which could potentially impact the results. Fively, this report represents a secondary analysis of existing databases. Although adjustments were made for many confounding factors, variables not included in the database were not adjusted for. However, we calculated the E value to quantify the potential impact of unmeasured confounders and found that unmeasured confounders were unlikely to explain the results. sixly, The current diagnostic criteria for prediabetes have some limitations, primarily relying on impaired fasting glucose and not considering oral glucose tolerance tests, glycated hemoglobin, and multiple fasting glucose measurements. This single-index approach may lead to underdiagnosis of prediabetes. Therefore, future studies evaluating the incidence of prediabetes should consider measuring additional variables, including oral glucose tolerance tests, glycated hemoglobin, and multiple fasting glucose measurements, to improve diagnostic accuracy and comprehensiveness. Additionally, this retrospective observational study provides an associative inference rather than establishing a causal relationship between the TyG index and the risk of prediabetes. Lastly, this study only measured the TyG index and other parameters at baseline and did not consider changes in the TyG index over time. Therefore, in the future, we should collect as much information as possible, including information on changes in the TyG index.
This study reveals a nonlinear relationship and a saturation effect between the TyG index and the risk of prediabetes among individuals aged 20 to 45 in China. There is a significant positive correlation between the TyG index and the risk of prediabetes to the left of the inflection point at a TyG index of 9.39. Reducing FBG and TG levels can significantly decrease the risk of prediabetes. Conversely, when the TyG index reaches or exceeds 9.39, its relationship with the risk of prediabetes is no longer statistically significant; merely lowering the TyG index is not enough to reduce the risk of prediabetes. It is also necessary to comprehensively control other factors such as BMI, smoking, and alcohol consumption. Understanding this non-linear relationship can help clinicians identify high-risk young individuals and implement focused interventions to reduce the risk of developing diabetes.
The author(s) declare financial support was received for the research, authorship, and/or publication of this article. The Guangdong Provincial Medical Science and Technology Research Fund project (NO: A2024429).
Ethics statement.
The studies involving humans were approved by The Rich Healthcare Group Review Board. The ethics committee waived the need for written informed consent. The studies were conducted in accordance with the local legislation and institutional requirements. Written informed consent for participation in this study was provided by the participants’ legal guardians/next of kin.
JX: Data curation, Writing – original draft, Writing – review & editing. LZ: Formal analysis, Writing – original draft, Writing – review & editing. CL: Data curation, Formal analysis, Writing – review & editing. YH: Formal analysis, Writing – review & editing. ZH: Formal analysis, Writing – original draft, Writing – review & editing.
The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.
All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.
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A cohort study is a type of observational study that follows a group of participants over a period of time, examining how certain factors (like exposure to a given risk factor) affect their health outcomes. The individuals in the cohort have a characteristic or lived experience in common, such as birth year or geographic area.
The cohort study design is an excellent method to understand an outcome or the natural history of a disease or condition in an identified study population (Mann, 2012; Song & Chung, 2010). Since participants do not have the outcome or disease at study entry, the temporal causality between exposure and outcome (s) can be assessed using this ...
Design, Analysis, and Reporting. Cohort studies are types of observational studies in which a cohort, or a group of individuals sharing some characteristic, are followed up over time, and outcomes are measured at one or more time points. Cohort studies can be classified as prospective or retrospective studies, and they have several advantages ...
The term "cohort" refers to a group of people who have been included in a study by an event that is based on the definition decided by the researcher. For example, a cohort of people born in Mumbai in the year 1980. This will be called a "birth cohort.". Another example of the cohort will be people who smoke.
Cohort Studies - Quantitative study designs
Abstract. Cohort studies are types of observational studies in which a cohort, or a group of individuals sharing some characteristic, are followed up over time, and outcomes are measured at one or more time points. Cohort studies can be classified as prospective or retrospective studies, and they have several advantages and disadvantages.
A cohort study is a particular form of longitudinal study that samples a cohort (a group of people who share a defining characteristic, typically those who experienced a common event in a selected period, ... failure to refute a hypothesis often strengthens confidence in it. Crucially, the cohort is identified before the appearance of the ...
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This design determines whether exposure to a risk factor affects an outcome. Cohort studies are a type of longitudinal study because they track the same set of subjects over time. For example, if researchers hypothesize that exposure to a chemical increases skin cancer, they can form a cohort based on exposure to that chemical.
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Background The health effects of Life's Essential 8 (LE8) on chronic diseases have been disclosed, but its association with hypertension remains unknown. The current study aimed to explore the potential link between 10-year LE8 trajectory and the incidence of hypertension. Methods LE8 was constructed from four behaviors and four metabolic factors, ranging from 0 to 100. Latent mixture models ...
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This population-based cohort study was based on the Swedish school-based HPV vaccination programme, which offered the first vaccine dose to girls aged 10-13 years (school grades 5-6), with a second dose offered within 12 months. 5 We identified all girls born between Jan 1, 2002, and March 1, 2004, using the Swedish Total Population ...
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A prospective cohort study conducted in rural China suggests that an increasing TyG index is associated with a higher cumulative risk of developing incident T2DM among ... consistent with our research hypothesis. That is for every 1-unit increase in the TyG index, the risk of prediabetes will increase by 81% (HR: 1.81, 95% CI 1.54-2.13, P < 0 ...