The initiation of the Canadian Multicentre Osteoporosis Study in 1996, and subsequent follow-up of the cohort 5 years later, provided longitudinal body mass index (BMI) data for a random sample of Canadians.Height and weight were measured at baseline and 5 years and used to calculate BMI and assign one of six weight categories. Multiple imputation was used to adjust for missing weight at year 5. Data were stratified by age and gender. The proportion of participants moving between categories was generated, and multivariable linear regression was used to identify factors associated with weight change.Baseline data were available for 8548 participants, year 5 data for 6721, and year 5 weight was imputed for 1827 (17.6%). Mean BMI for every age and gender group exceeded healthy weight guidelines. Most remained within their BMI classification over 5 years, but when change occurred, BMI category was more likely to increase than decrease. Several sociodemographic, lifestyle and clinical characteristics were associated with change.Mean baseline BMI tended to be higher than recommended. Moreover, on average, men under age 45 and women under age 55 were gaining approximately 0.45 kilograms (one pound) per year, which leveled off with increased age and reversed in the oldest age groups. These findings underscore the need for public health efforts aimed at combating obesity. The World Health Organization has recommended an updated classification system for the assessment of body mass index  in adults . This syIt is widely accepted that given a certain height, there is a range of weight that is associated with good general health ,3. DespiBeing overweight or obese can have a significant impact on health, as these individuals are more likely to suffer from a variety of illnesses ,10,11,13Few longitudinal studies have assessed changes in BMI over time. Data with measurements at two-year intervals from the Framingham Heart Study suggest that BMI tends to be stable over time . The Canos) in 1996, and subsequent follow-up of the cohort 5 years later, provided this opportunity. CaMos is not a study of individuals with osteoporosis; rather, it was designed to study the incidence and prevalence of osteoporosis in a random sample of Canadians 25 years of age and older [There is a need for a population-based study of measured BMI in order to examine longitudinal weight change ,7,13. Thnd older . This alos is an on-going, prospective cohort study of 9,423 non-institutionalized, randomly selected men and women aged 25 years and older at baseline, drawn from a 50-kilometer radius of nine Canadian cities . Baseline assessments took place between February 1996 and September 1997, and the 5-year follow-up took place between February 2001 and September 2002. All data were obtained by trained interviewers, who used a questionnaire with good reproducibility that was designed for the CaMos study. Participants provided written consent.CaMos is available elsewhere [os was designed to collect epidemiological data related to the incidence and prevalence of osteoporosis, so although the sampling framework was random, it was designed to include more women than men, and a higher number of older than younger Canadian residents.A detailed description of the objectives, methodology and sampling framework for CaMlsewhere . BrieflyEthics approval was obtained through the Review Boards of each participating centre. These included Memorial University of Newfoundland Human Investigation Committee (St John's), the Capital Health Research Ethics Board , Comité d'éthique de la recherché clinique du Centre Hopitalier Universitaire de Québec (Quebec City), St. Michael's Hospital Research Ethics Board (Toronto), McMaster University Research Ethics Board (Hamilton), Queen's University Health Sciences and Affiliated Teaching Hospitals Research Ethics Board (Kingston), University of Saskatchewan Biomedical Research Ethics Board (Saskatoon), University of Calgary Health Research Ethics Board , University of British Columbia Clinical Research Ethics Board (Vancouver) and the McGill University Health Centre Research Ethics Committee .The majority of participants were scheduled for dual-energy x-ray absorptiometry (DXA) assessment of bone mineral density (BMD) as part of the study protocol, at which time both height and weight were measured. Height was measured without shoes, using a wall-mounted stadiometer, ruler on the wall or a measure on a weight scale. Weight was measured in light clothing using a beam balance or electronic scale. For those who elected not to have the DXA, or for whom it could not be scheduled, the interviewer measured height and weight with a carpenter's rule and a portable scale. Baseline height and weight were used to calculate the baseline BMI. Five-year BMI was calculated using the baseline height and the weight at year 5. Baseline height was used to avoid the need to impute height for those missing height at year 5, as the loss of height over time was expected to be close to zero for the majority of participants. Change in BMI was calculated as year 5 minus the baseline value, so that positive values would indicate an increase in weight.Multivariable linear regression was used to identify the factors associated with five year changes in BMI for men and women. Variables were identified on the basis of an extensive literature review, and included sociodemographic characteristics , clinical characteristics , and other factors . In addition, since we expected that changes in some of these variables could be associated with changes in BMI, change variables were also examined for menopausal status, comorbidities, smoking, alcohol use, the activity variables, level of pain, level of happiness, and perception of general health.a priori considerations  were retained due to widespread belief that the variable was associated with weight change [Univariate associations between all variables and change in BMI were examined to refine the list of predictors identified above, removing any for which there was no evidence to support its inclusion. For some, the 95% confidence interval (CI) may have included zero but if one or both of the upper or lower limits could be of clinical interest, the variable was retained, allowing the investigators to see the effects of all variables of potential interest. Three variables not automatically retained in the model by these t change ,24-26.Not all participants who provided baseline data participated at year 5, and differences in the characteristics between participants and non-participants could cause the estimates to be biased. For example, this could occur if subjects with certain diseases tended to lose weight or not participate because they were unwell. Multiple imputation  was therWithin the multiple imputation models, linear regression models were fit to predict weight at year 5, using data from respondents with complete data at both time points. Models included baseline height and weight, age, and BMD of the femoral neck from year 5 if available, otherwise at baseline. Separate models were developed for men and women within each age group. Covariate data from those missing year 5 weights were then entered into the models to impute the missing weights. Simultaneous regression parameter estimations and multiple imputations were carried out via a Gibbs sampler algorithm as implemented in WinBugs software .Complete baseline data were available for 2572 men and 5976 women, for a total of 8548 . Height and/or weight were missing for 277 at baseline, primarily due to scheduling difficulties for the DXA at two sites. Six were pregnant at baseline or at follow-up. Six hundred and forty-four had died by the time of the 5-year follow-up, and although 592 of these had provided baseline data , their data were not used since 5-year BMI changes were undefined for these subjects.At year 5, data were available for 1959 (76.2%) men and 4762 (79.7%) women, for a total of 6721, and were imputed for 1827 (613 men and 1214 women). Of these 1827, 764 were still in the cohort but were missing weight at year 5, largely due to scheduling difficulties and one site that had not measured weight at the time of the DXA. Another 415 completed a short questionnaire over the telephone. The reasons for loss to follow-up included 265 who could not be contacted, 106 who withdrew consent, 89 who were too sick, 114 who had moved away, and 74 with miscellaneous reasons such as cancelled, no time, or no reason given.Those without complete year 5 were represented in all categories of weight. For men, 33.3% of those in the underweight group and 35.0% of those in the obese class III group at baseline were missing year 5 data. For other weight categories, the proportions with incomplete year 5 data ranged from 21.4% (overweight) to 26.3% (obese class I). Among women, the proportions with incomplete data ranged from 19.1% of those in the normal weight category to 27.8% of those in obese class III. For the other categories, the proportions were very similar, ranging from 20.5% (overweight) to 21.8 (obese class II), suggesting that there was no bias due to greater loss to follow-up within certain weight categories.1,2. The lower limit of the 95% CI was less than 25 for only one group (women 25–34 years), and was exactly 25.0 for women aged 35–44 years. This figure also contains the mean 5-year change in BMI with the 95% CI for the age and gender groups. The increases in BMI were much larger for men between 25 and 44 years and women between 25 and 54 years than in the older groups, in which the BMI tended to decrease.Figure Tables Tables The results of movement from one BMI category to another over 5 years are very similar for women Table . A womanTable Current smokers and those who had never smoked tended to gain less weight as compared to those who had quit smoking, although only one of the four comparisons attained clinical importance in that the 95% CI did not include zero. Higher alcohol consumption was associated with small declines for both men and women, but the results were inconclusive for both. Those who felt less happy or whose happiness assessment stayed about the same, also saw less gain as compared to those who indicated that their level of happiness had improved, although this result was inconclusive for men. Some categories of changes in participation in regular activity, change in number of sedentary hours, and perception of overall health were also associated with increases in BMI for either men or women, although not always in the expected direction. For example, those who indicated that their number of sedentary hours was quite low still saw a small increase in the parameter estimates.For women, the association between number of live births and weight change was virtually non-existent. Finally, after adjusting for age, the effect of menopause was also inconclusive, although it is of clinical interest that those who had not yet become menopausal, or who had become menopausal recently, saw small declines as compared to those who were already menopausal at baseline.These results add to the growing body of literature suggesting that the prevalence of overweight and obesity is increasing. The mean baseline BMI exceeded 25.0 for every age and gender category studied, ranging from 25.6 (women aged 35–44) to 27.7 (men aged 55–64). The mean change over 5 years does not appear large at face value, ranging from an increase of 1.07 for the youngest group of men to a drop of 0.96 for the oldest group of women. However, if one adds the mean change to the mean baseline value, all values remain greater than 25, and the values increase for all but the two oldest age groups, for both men and women. In view of the fact that the mean BMI exceeded 25 even in the youngest age groups in both men and women  would weigh 49.1 kg for a BMI of 18.50 and 66.3 kg for a BMI of 25.0, representing an increase of 17.2 kg (37.9 pounds) to move into the overweight category if starting at the lowest point in the normal weight category. To do the same, a man at the average Canadian height of 178 cm would weigh 58.5 kg for a BMI of 18.50 and 79.1 kg for a BMI of 25.0, representing an increase of 20.6 kg (45.4 pounds). In these two examples, to move up by one BMI point, the woman would have to gain 2.6 kg (5.7 pounds), while the man would have to gain 3.1 kg (6.8 pounds). Given the results in Figure The regression models suggest that there are factors associated with change in BMI, but much of the variance remains unexplained, and the models must be interpreted with caution due to the wide confidence intervals associated with many of the estimates. Nevertheless, the findings tend to be consistent with previous research, although there are some notable exceptions. For example, several studies have noted an association between increased weight and a variety of comorbid conditions ,11 as weOur data did suggest a moderate effect of region, with the central region most likely to show an increase in BMI. One previous Canadian study noted almost no effect of region , but a sCurrent smokers had less gain than those who had never smoked or had recently quit, particularly for women, which is consistent with other research ,22, whilAll education levels below the University level were associated with more weight gain, which was more pronounced for women and consistent with past findings ,10,18. NThe ability to assess the effect of behaviour change, and its impact on change in weight, is one of the strengths of this study. Compared to those who did not participate in physical activity but do now, all other groups showed small BMI gains, supporting other research that demonstrated a healthy effect of increased activity on BMI ,29. MoreOne limitation to this study was the loss to follow-up, as only 90.7% of the baseline sample could be included even when using multiple imputation. Moreover, multiple imputation also has assumptions and limitations. For example, the technique assumes that the baseline data were sufficiently detailed to predict year 5 weights, and that, given the information available for predicting missing year 5 weights, those who participated are similar at baseline to those who did not participate . It is pos cohort.Caution must be used when interpreting any results based on BMI data. While BMI is a commonly used indicator of weight category, it is a composite measure that is unable to distinguish between fat and lean tissue . MoreoveThis study provides evidence that Canadians have a higher BMI than that which is currently recommended for optimal health. Moreover, weight appears to be increasing in younger age groups, although this reverses after age 65 in both men and women. These findings suggest that a substantial number of people may be putting their health at risk due to being overweight or obese. Although there is a tendency to stay in the same weight category, this increase in BMI over time underscores the need for public health efforts aimed at combating the current epidemic of obesity, especially in younger men and women.The author(s) declare that they have no competing interests.WMH, CB, SIB, JCP, MH and TT participated in the conception and design of this CaMos ancillary study; CL, CB and LJ were responsible for the analysis; all were involved in the interpretation of the findings and drafting the manuscript; all have read and approved the final manuscript.os Coordinating Centre, McGill University, Montreal, Quebec: David Goltzman , Alan Tenenhouse , Suzette Poliquin , Suzanne Godmaire (research assistant), Claudie Berger (study statistician), Lawrence Joseph (consultant statistician).CaMMemorial University, St. John's Newfoundland: Carol Joyce (director), Christopher Kovacs (co-director), Emma Sheppard (coordinator).Dalhousie University, Halifax, Nova Scotia: Susan Kirkland, Stephanie Kaiser (co-directors), Barbara Stanfield (coordinator).Laval University, Quebec City, Quebec: Jacques P. Brown (director), Louis Bessette (co-director), Marc Gendreau (coordinator).Queen's University, Kingston, Ontario: Tassos Anastassiades (director), Tanveer Towheed (co-director), Barbara Matthews (coordinator).University of Toronto, Toronto, Ontario: Bob Josse (director), Tim Murray (co-director), Barbara Gardner-Bray (coordinator), Nancy Kreiger .McMaster University, Hamilton, Ontario: Jonathan D. Adachi (director), Alexandra Papaioannou (co-director), Laura Pickard (coordinator).University of Saskatchewan, Saskatoon, Saskatchewan: Wojciech P. Olszynski (director), K. Shawn Davison (co-director), Jola Thingvold (coordinator).University of Calgary, Calgary, Alberta: David A. Hanley (director), Jane Allan (coordinator).University British Columbia, Vancouver, British Columbia: Jerilynn C. Prior (director), Yvette Vigna (coordinator).The pre-publication history for this paper can be accessed here:


The Alzheimer's Disease Anti-inflammatory Prevention Trial (ADAPT) was designed to evaluate the conventional NSAID naproxen sodium and the selective COX-2 inhibitor celecoxib for primary prevention of Alzheimer's dementia (AD). On 17 December 2004, after the Adenoma Prevention with Celecoxib (APC) trial reported increased cardiovascular risks with celecoxib, the ADAPT Steering Committee suspended treatment and enrollment. This paper reports on cardiovascular and cerebrovascular events in ADAPT.ADAPT is a randomized, placebo-controlled, parallel chemoprevention trial with 1–46 mo of follow-up.The trial was conducted at six field sites in the United States: Baltimore, Maryland; Boston, Massachusetts; Rochester, New York; Seattle, Washington; Sun City, Arizona; and Tampa, Florida.The 2,528 participants were aged 70 y and older with a family history of AD.Study treatments were celecoxib (200 mg b.i.d.), naproxen sodium (220 mg b.i.d.), and placebo.Outcome measures were deaths, along with nonfatal myocardial infarction (MI), stroke, congestive heart failure (CHF), transient ischemic attack (TIA), and antihypertensive treatment recorded from structured interviews at scheduled intervals. Cox proportional hazards regression was used to analyze these events individually and in several composites.Counts (with 3-y incidence) of participants who experienced cardiovascular or cerebrovascular death, MI, stroke, CHF, or TIA in the celecoxib-, naproxen-, and placebo-treated groups were 28/717 (5.54%), 40/713 (8.25%), and 37/1070 (5.68%), respectively. This yielded a hazard ratio  for celecoxib of 1.10 (0.67–1.79) and for naproxen of 1.63 (1.04–2.55). Antihypertensive treatment was initiated in 160/440 (47.43%), 147/427 (45.00%), and 164/644 (34.08%). This yielded hazard ratios (CIs) of 1.56 for celecoxib (1.26–1.94) and 1.40 for naproxen (1.12–1.75).For celecoxib, ADAPT data do not show the same level of risk as those of the APC trial. The data for naproxen, although not definitive, are suggestive of increased cardiovascular and cerebrovascular risk. Background: Evidence from observational studies suggests that people taking certain nonsteroidal anti-inflammatory drugs (NSAIDs) are at lower risk of developing Alzheimer's disease. However, in order to reliably find out whether NSAIDs reduce the risk of Alzheimer's, it is important to perform a properly designed randomized trial. Such a trial, ADAPT, was sponsored by the United States National Institute on Aging, and the study started recruitment in 2001. The trial involved three treatment arms: naproxen (one type of NSAID), celecoxib , and placebo, acting as a control. It was planned that 2,625 participants would be recruited and that the primary outcome of interest was incidence of Alzheimer's disease in the three treatment arms; the trial would run for 7 y. However, this trial was terminated early, a decision based in part on information from other studies that demonstrated an increased risk of certain harms, such as heart attacks and strokes, in people taking celecoxib and other types of COX-2 inhibitors. Therefore meaningful data were not available at the time on the study's primary outcome (prevention of Alzheimer's disease). However, data about the chance of these harms are available from the ADAPT results, and these results are presented here.What this trial shows: The investigators compared frequency of particular types of harm in the treatment arms: heart attack, stroke, congestive heart failure (CHF), and transient ischemic attack (TIA). For each individual type of event, some were more likely in people treated with celecoxib compared with placebo, but others were not. When considering people taking naproxen, all four types of adverse events were more likely to occur in the treatment group as compared to placebo. The investigators then combined data from all four types of harm together, and here they found that the overall risk in people taking celecoxib was higher than for people taking placebo, but that this was not statistically significant, so it could have been due to chance alone. When considering naproxen as compared with placebo, the researchers saw an approximately 60% increase in risk for all four harms combined, and this result was statistically significant. The death rate in people taking either celecoxib or naproxen was higher than for those taking placebo, but this was not statistically significant, and therefore could have been due to chance.Strengths and limitations: Strengths of this study include the randomization procedures, which used a distributed computer system to assign patients to treatment arms (minimizing the chance of bias), blinding of patients to their treatment assignment, and blinding of the committee reviewing deaths and safety reports to treatment assignment. One limitation is that although the trial was large and appropriately powered for the main outcome (prevention of Alzheimer's disease), the number of safety events reported here were small and the trial was not primarily designed to examine safety. Further, participants eligible to join this trial were required to have a family history of Alzheimer's disease, so it is possible that their risk factors are slightly different from the general population.Contribution to the evidence: The cardiovascular safety of NSAID's, including COX-2 inhibitors, is an intensely debated topic. Very few published data exist on the long-term safety of celecoxib as compared with placebo, although there are a number of as-yet-unpublished studies. These data on harms provided by ADAPT provide important results that should be incorporated into future meta-analyses. Such meta-analyses will give a more rigorous and reliable assessment of the safety of the drugs studied here. Epidemiological studies suggest that nonsteroidal anti-inflammatory drugs (NSAIDs) may delay or prevent the onset of Alzheimer's dementia (AD). A recent meta-analysis of three prospective studies indicated a relative hazard  of 0.42 (0.26–0.66) among individuals with 2 y or more of sustained NSAID treatment, compared to nonusers . The AlzOn Friday, 17 December 2004, increased cardiovascular risks with celecoxib were reported from the National Cancer Institute-sponsored Adenoma Prevention with Celecoxib (APC) trial. On the same day, the ADAPT Steering Committee suspended treatment with celecoxib as well as with naproxen. Suspension of naproxen reflected the ADAPT investigators' reluctance to imply, by continuing the trial, that naproxen was safer than celecoxib when ADAPT data did not support this conclusion. Our rationale for suspending both treatments in ADAPT was more fully discussed in our 18 February 2005 testimony to the Food and Drug Administration (FDA) Arthritis Advisory Committee and the Drug Safety and Risk Management Advisory Committee [This report describes the structure of ADAPT and presents the trial's safety data through the end of the treatment period. We offer our interpretation of these data in their context, that is, for a trial that was not designed or powered for cardiovascular or cerebrovascular events as primary outcomes and, additionally, was stopped early. We believe we are obligated to make these results available to the scientific community and the public at large, and we present them as a contribution to the larger body of evidence about the potential cardiovascular and cerebrovascular risks of COX-2 inhibitors and conventional NSAIDs.ADAPT is a randomized, placebo-controlled, multicenter, primary prevention trial. It is investigator-initiated and sponsored by the National Institute on Aging (NIA) via a cooperative agreement. All pertinent institutional review boards approved the study protocol.http://www.jhucct.com/adapt/manall43.pdf). Written consent was obtained from each participant.Recruitment was accomplished primarily through mailings to Medicare beneficiaries targeted by age and by ZIP code to areas surrounding the trial's six field sites in the United States . Eligibility criteria included being the age of 70 y or older and having a history of at least one first-degree relative with Alzheimer-like dementia. Participants underwent cognitive screening to exclude those with dementia or other cognitive disorders. Persons regularly using NSAIDs were excluded. Use of aspirin was allowed at 81 mg per day or less. More specific information on eligibility criteria is available from the trial's protocol (available at Participants were randomly assigned to receive celecoxib (200 mg b.i.d.), naproxen sodium (220 mg b.i.d.), or placebo.The randomization sequence was generated by the trial's coordinating center in permuted blocks stratified by three age groups  and the six field sites, with an assignment ratio of 1:1:1.5.Treatment assignment was in the form of an allocation to one of 49 bins . The bin system simplified long-term supply of study drug, and the number of bins reduced the risk of unmasking large numbers of participants. Only select coordinating center personnel knew the correspondence of bin allocation to treatment assignment.Randomization was implemented via computer systems distributed to each of the six sites. The sequence of bin assignments was concealed from site personnel via encoding and password protection of the randomization files. Before randomization, site personnel were required to enter baseline data. The ADAPT computer system confirmed eligibility before releasing the bin assignment.Masking of participants and field site personnel was achieved using a double placebo design ; each biSample size calculations were performed using a SAS macro by Shih  for timeStudy clinicians collected data through in-person interviews of participants 1 mo and 6 mo after randomization and then every 6 mo thereafter. Similarly constructed telephone interviews with participants were conducted every 6 mo in the intervals between in-person visits, starting at month 3. During these routine visits and calls, masked study personnel recorded participant reports of adverse events on data forms. When indicated, according to clinical judgment, they clarified these self-reported events with participants' primary care physicians. All fatal events also were reported in real time on death reports, and death certificates were obtained when possible.The outcomes we report here are adverse events, including occurrences of death, myocardial infarction (MI), stroke, congestive heart failure (CHF), and transient ischemic attack (TIA). Fatal events were obtained from death reports, and nonfatal events were determined from scheduled follow-up interviews. Deaths are included if they occurred on or before 17 January 2005 (1 mo after suspension of treatment) and were reported by 1 July 2005. An internal committee of three physicians, who were masked to treatment assignment, reviewed all events with details recorded on death reports to classify the fatal events. Events reported during scheduled follow-up contacts are included if they were recorded by 17 April 2005  and keyed into the ADAPT database by 1 July 2005. Composite outcomes were created in a manner similar to that of the APC trial  .We also analyzed new occurrences of treatment for hypertension among participants not receiving treatment for hypertension at baseline. This information was obtained by participant report at routine data collection times.p-values. Analyses of deaths included all participants, with censoring during follow-up of individuals with unknown vital status. Analyses of events ascertained at scheduled follow-up excluded participants with no follow-up data collection and censored those lost to follow-up. Active treatments were compared individually with placebo, but not with one another. The Kaplan-Meier method was used to graph the incidence of the composite outcome. Secondary analyses considered events occurring while participants were being issued the study drug, and events were stratified by baseline use of aspirin for cardioprotection. All analyses were conducted with SAS version 8.1.Percentages of participants experiencing the events of interest were calculated by assigned treatments. Analyses of time to first occurrences of events were performed using Cox proportional hazards regression, and hazard ratios were estimated with 95%CI and Wald The trial's Treatment Effects Monitoring Committee met in person twice a year to review efficacy and safety data classified by treatment assignment. The committee, comprising four independent voting members and three nonvoting members from the trial , was advisory to the Steering Committee and the NIA.The flow of participants through randomization, follow-up, and analysis is diagramed in p = 0.32).All participants contributed to analyses of mortality. Vital status was unknown for only three participants, who could not be contacted at the time of treatment suspension or within a 6-mo window before or after that time. Of the participants, 2,500 (98.9%) had follow-up for nonfatal events. Participants contributed 4,660 person-years of follow-up through 17 January 2005 . The median follow-up times for the three treatment groups were 23.3, 23.5, and 22.1 mo, respectively. Median times of administration of the study drug were 16.1, 14.0, and 15.6 mo, respectively (p = 0.56 for celecoxib and p = 0.59 for naproxen).Hazard ratios were similar for secondary analyses that considered only the time during which participants were administered study treatments. Analyses stratified for baseline use of aspirin showed the highest risk among those in the naproxen group who were not taking aspirin, but the tests of interaction by aspirin use were not significant for either drug  at entry and the low mortality rate observed during the trial. However, these factors limit generalizability only to the extent that they modify the treatment effect. There has been speculation that any increased cardiovascular risk from anti-inflammatory drugs may be confined to those at higher risk of disease, but there is little rationale for or evidence of differing response to anti-inflammatory treatment.The cardiovascular safety of COX-2 inhibitors has been questioned since the publication of the Vioxx Gastrointestinal Outcomes Research (VIGOR) trial results in late 2000, in which rofecoxib (Vioxx) was associated with a 5-fold increased risk of MI . The CelOn 30 September 2004 Merck withdrew rofecoxib from the market, after results from the Adenomatous Polyp Prevention on Vioxx (APPROVe) trial showed a statistically significant relative risk for thrombotic events of 1.92 for rofecoxib compared with placebo . On 15 OAs of the writing of this manuscript, the APC paper  is the oWhile the ADAPT celecoxib data may be compared with corresponding results from sister trials of similar size and duration, our naproxen data stand alone. Several trials have reported that naproxen produced lower cardiovascular risks than comparator COX-2 inhibitor treatments, and the meta-analysis by Kearney et al.  estimateIt has been suggested that naproxen, at least at higher doses, produces lower cardiovascular risk than comparator COX-2 drugs because of antithrombotic effects similar to those of aspirin. Numerous observational studies have investigated the association of naproxen use and CVD, and a meta-analysis of eight case-control studies and three retrospective cohort studies obtained a combined estimate of 0.86 . HoweverSeveral published trials tested naproxen and placebo as well as a COX-2 inhibitor, and three reported on the outcome of incident hypertension –30. One The ADAPT data on naproxen are by no means definitive. Under ordinary circumstances, they would not have warranted suspension of treatment in December 2004. Nonetheless, after the APC announcements, the ADAPT investigators were fundamentally uncomfortable with continuing naproxen simply because highly publicized external circumstances had spotlighted risks with celecoxib, but not with naproxen. It has been suggested that all NSAIDs may have some degree of cardiovascular risk ,26,31, pThere are several limitations in the ADAPT data. First, because ADAPT was not designed to detect differences in cardiovascular and cerebrovascular risks, the composite outcomes were specified post hoc. We designed their constituent events both to be as comparable as possible to the approaches used in the Antiplatelet Trialists Collaboration and the APC trials and to draw on outcomes collected in a systematic manner from close-ended questions asked to participants at every follow-up contact. Because ADAPT by design focused on neurological events, we asked participants specifically about the occurrence of both stroke and TIA. Although we also asked about MI and CHF, we did not ask specific questions about conditions such as unstable angina or other indications for interventional cardiologic procedures and therefore were not able to include them in our composite of systematically collected outcomes.We also did not have a priori procedures for adjudication of cardiovascular or cerebrovascular events. Although we could employ such procedures post hoc in an attempt to increase the precision of reported events, these procedures have several limitations. We cannot identify missed events, and we cannot be certain that the process of adjudication is not influenced by the current controversy. Therefore, adjudication will only decrease the absolute risk observed and could bias the relative risk. Also, adjudication may foster a sense of precision that is inappropriate given the small numbers of reported events in a trial that was terminated early . For theUltimately, the largest limitation in the ADAPT data is the imprecision resulting from the small numbers of events. A corresponding caution in their interpretation is warranted. Even so, our actions, as well as those of other investigators and the FDA, reflect the low tolerance for risk in primary prevention trials. For their participants, the only benefit of the intervention is the possibility of later protection from a condition they might have avoided in any case. This creates some difficulty for extrapolation to the clinical setting, where these drugs are used for pain relief in conditions such as osteoarthritis and rheumatoid arthritis, and the potential risks need to be weighed against the known immediate benefits of treatment. Nevertheless, we as a medical and research community are attempting just such difficult extrapolation, as the primary prevention setting is providing long-term, placebo-controlled data for drugs approved on the basis of short-term trials but used for the treatment of chronic conditions frequently necessitating decades of analgesic therapy. Therefore, we believe it is essential to make data from long-term trials such as ADAPT available to contribute to the larger body of evidence on drug safety.CONSORT ChecklistClick here for additional data file.(47 KB DOC)Trial Protocol: ADAPT Protocol, Version 1.4Click here for additional data file.(233 KB PDF)The writing committee that takes primary responsibility for this paper (see Acknowledgments) reports the following competing interests. Barbara Martin, Denis Evans, and Curtis L. Meinert have no conflicts of interest to report. John C. S. Breitner previously held partial royalty interests in two United States patents for the use of NSAIDs for the prevention of Alzheimer's disease; in January 2005 he assigned these interests irrevocably as a gift to a private charitable foundation in which he has no constructive or beneficial interest. Constantine G. Lyketsos is a consultant to GlaxoSmithKline and receives lecture honoraria and support for continuing medical education activities from AstraZeneca, Eisai, Forest Laboratories, Janssen, Lilly, Novaris, and Pfizer.Along with the competing interests listed by the writing committee, the Steering Committee's competing interests are as follows. Drs. Mullan and Piantadosi have no conflicts of interest to report. Dr. Green has received lecture honoraria and support for continuing medical education activities from Eisai, Forest Laboratories, and Pfizer and is an unpaid scientific advisor to Myriad Pharmaceuticals. Dr. Tariot is a consultant to Abbott Laboratories, AstraZeneca, Bristol-Meyers Squibb, Eisai, Forest Laboratories, Janssen, Lilly, Myriad Pharmaceuticals, Pfizer, Sanofi, and Schwabe; he has received research support from Abbott Laboratories, AstraZeneca, Elan, Pfizer, Bristol-Myers Squibb, Forest Laboratories, Merck, Neurochem, Ono, and Sanofi and lecture honoraria from Pfizer, Forest Laboratories, AstraZeneca, and Eisai. Dr. Craft is a consultant for and receives research support from GlaxoSmithKline. Dr. Sabbagh is a speaker for Pfizer, Eisai, Ortho-McNeil, Forest Laboratories, and Abbott Laboratories and receives research support from GlaxoSmithKline, Pfizer, Eisai, Elan, Ono, and Takeda. Dr. Brandt receives royalties from Psychological Assessment Resources, Inc., publisher of the Hopkins Verbal Learning Test—Revised, one of the neuropsychological tests used in this trial; however, Dr. Brandt waived royalties from the use of the test in ADAPT. Dr. Welsh-Bohmer, who is a non-voting member of the Steering Committee, continues to hold royalty interests in these two patents; she also is a paid consultant to GlaxoSmithKline and has received speaking honoraria from Pfizer and private gifts to the Ryan ADRC from GlaxoSmithKline and Novartis.


We estimated the median and 95th percentile of daily exposures to DBP to be 0.99 and 2.68 μg/kg/day, respectively; for DEP, 6.64 and 112.3 μg/kg/day; for BBzP, 0.50 and 2.47 μg/kg/day; and for DEHP, 1.32 and 9.32 μg/kg/day. The U.S. Environmental Protection Agency (EPA) reference doses for these chemicals are 100 (DBP), 800 (DEP), 200 (BBzP), and 20 (DEHP) μg/kg/day. The median and 95th percentile exposure estimates for the phthalates associated with reduced anogenital distance in the study population are substantially lower than current U.S. EPA reference doses for these chemicals and could be informative to any updates of the hazard assessments and risk assessments for these chemicals.Phthalate diesters have been shown to be developmental and reproductive toxicants in animal studies. A recent epidemiologic study showed certain phthalates to be significantly associated with reduced anogenital distance in human male infants, the first evidence of subtle developmental effects in human male infants exposed prenatally to phthalates. We used two previously published methods to estimate the daily phthalate exposures for the four phthalates whose urinary metabolites were statistically significantly associated with developmental effects in the 214 mother–infant pairs [di- Phthalates are used in a variety of industries and are present in many consumer products, such as soaps, perfumes, cosmetics, shampoos, building products, shower curtains, aerosols, plastic toys, and plastic packaging , 2003. Dn-butyl phthalate , and monoisobutyl phthalate  were significantly associated with reduced AGD and anogenital index (AGI = AGD/body weight) in male infants.In a study published in a previous issue of this journal by some authors participating in the current study, Although none of the 134 boys examined showed frank malformations or disease, and 86.6% of these boys had both testicles classified as normal, AGI was significantly correlated with degree of testicular descent as well as penile volume and scrotal size . The medThe current U.S. Environmental Protection Agency (EPA) reference doses (RfDs) for DEP, DBP, and BBzP were formulated in the early 1990s using older animal studies  , 2005d. In this study, we applied a simple pharmacokinetic model, initially proposed by Women included in this study were originally recruited into the Study for Future Families (SFFI), a multicenter pregnancy cohort study, at prenatal clinics in Los Angeles, California (Harbor-UCLA and Cedars-Sinai); Minneapolis, Minnesota ; and Columbia, Missouri (University Physicians), from September 1999 through August 2002. Details of study participation are given by n = 214) are shown in n = 85) used by In the We calculated daily exposure for the phthalate metabolites that were statistically significantly associated with reduced AGI in the FE is the total fraction and FU is the urinary fraction of the dose eliminated in time t, and ktotal and ku are the apparent first-order rate constants for total elimination and urinary elimination of monoester, respectively. We calculated the two rate constants, ktotal and ku, by using previously published values for the excreted fractions of each parent diester , and MWd and MWm are the molecular weights of the diesters and monoesters, respectively. We used a value of 18 mg/kg/day for CE (ME) for each subject in the study. The unadjusted and creatinine adjusted phthalate urinary concentrations from the 214 samples from the where y for CE  and creaFor comparison, we also estimated the daily exposure using a second formula published by FUE, the molar fraction of the urinary excreted monoester related to the parent diester, is used in place of f. The fractional urinary excretion values for DBP (0.69) and BBzP (0.73) were taken from published human data . DEHP is initially metabolized to MEHP, which is then further metabolized to various other products, including MEHHP and MEOHP. All three metabolites are thought to be toxic , and estd as MBP .The results for the exposures of DEP, DBP, and BBzP of the women in the The estimated exposures calculated using the We have estimated exposures to a variety of phthalate diesters in a population of mother–infant pairs in which subtle developmental effects were associated with prenatal urinary phthalate metabolite concentrations. The models we have used to estimate exposures make no assumptions regarding the route of exposure. There are multiple possible routes of exposure to phthalates, including dermal , ingestin = 214) are on the order of 1 μg/kg/day, and for DEP are on the order of 6 μg/kg/day. Current U.S. EPA RfDs are 100 μg/kg/day (DBP), 200 μg/kg/day (BBzP), and 800 μg/kg/day (DEP), which are all more than 100 times greater than the median exposures in the The median estimated exposures for DBP and BBzP in the FE and FU). Because Kohn et al. used animal excretion data for some of the metabolites, they estimated that their FE values were accurate to approximately 50%, whereas the FU values could vary by 15-fold among species with humans in the middle. However, we used fractional urinary excretion values obtained from human studies in our calculations using the David formula. There has been much scientific debate regarding the appropriate use of FUE values when using the David formula to calculate DEHP exposure values. FUE for MEHP of 13%, calculated from human excretion data provided by FUE for MEHP of 2.4% and also provide a mathematical argument against the feasibility of 13% as the FUE for MEHP. The choice of FUE values is important because it affects the results of the exposure calculations. We use FUE values from the most recent human excretion data on DEHP , and with our calculations using the Kohn et al. method, which does not use FUE values. Our exposure estimates from the Kohn et al. and David formulas are similar, suggesting reasonable agreement between the models and parameters used.There are potential sources of uncertainty in the  on DEHP . Our MEHThe exposures within this study population of pregnant women are similar to or somewhat lower than those documented in other populations of women of reproductive age. Median female  MBP concentrations in NHANES 2001–2002 (21.5 μg/g creatinine) compare closely with those in the Except for the As discussed in the In addition, the observed associations in the


Furthermore, estrogen confers atheroprotection via COX-2–dependent PGI2 in mice, raising the possibility that COX inhibitors may undermine the cardioprotection, suggested by observational studies, of endogenous or exogenous estrogens.Suppression of prostacyclin (PGIp < 0.002) interaction. The OR when taking aspirin at doses of 150 mg/d or more was 1.41 (95% CI 0.47–4.22). However, a similar interaction was not observed with other commonly used drugs, including lower doses of aspirin, which target preferentially COX-1.To identify an interaction between hormone therapy (HT) and COX inhibition, we measured a priori the association between concomitant nonsteroidal anti-inflammatory drugs (NSAIDs), excluding aspirin, in peri- and postmenopausal women on HT and the incidence of myocardial infarction (MI) in a population-based epidemiological study. The odds ratio (OR) of MI in 1,673 individuals and 7,005 controls was increased from 0.66  when taking HT in the absence of traditional (t)NSAIDs to 1.50 (95% CI 0.85–2.64) when taking the combination of HT and tNSAIDs, resulting in a significant  consisting of replacement estrogen, and sometimes progesterone as well, is often taken by women experiencing symptoms of menopause. Evidence from observational studies and the Womens' Health Initiative (WHI) trial has suggested that HT protects against heart disease in perimenopausal women. However, researchers have suggested that any beneficial effect of hormone replacement therapy on the heart might be counteracted by the effects of certain types of painkillers also being taken by women involved in the studies. These painkillers, nonsteroidal anti inflammatory drugs ( NSAIDs), prevent production of a molecule called prostacyclin. Prostacyclin plays a role in preventing blood clotting and is therefore thought to be important in protecting the heart. Estrogen, however, acts to increase production of prostacyclin, and it is therefore theoretically possible that hormone replacement therapy does have a beneficial effect on heart health, but which is counteracted by the negative effects of NSAIDs.In this study, the researchers wanted to find out whether there was any evidence for an interaction between NSAID use, hormone replacement therapy, and heart disease. Such understanding in turn might help to identify more clearly whether hormone replacement therapy protects against heart disease in specific subgroups of postmenopausal women.This study was carried out using information from the UK's General Practice Research Database, which is the largest computer database of anonymous medical records from primary care anywhere in the world. It contains information entered by UK general practitioners on their patients' drug prescriptions, diagnoses, referrals to hospital, and other data. The researchers here searched for all individuals from the database who were aged between 50 and 84 years on 1 January 1997, and then followed them up through the database for four years, or until the individual died, reached 85 years of age, or was diagnosed with a heart attack or cancer. From this search, the researchers found 1,673 women who had heart attacks or who died from coronary heart disease; these were considered “cases.” Then, these 1,673 women were matched against 20,000 “control” women of similar age. Information was pulled out for each case or control on their use of hormone replacement therapy, NSAIDs , and various risk factors for heart disease. The researchers then compared use of hormone replacement therapy and NSAIDs between the cases and controls, while making statistical adjustments for other risk factors .   The researchers found that current use of hormone replacement therapy was associated with a lower risk of heart attack than non-use. The odds ratio (chance of a heart attack among HT users compared to the chance among non-users of HT) was 0.78. However, when looking at women who used NSAIDs at the same time as hormone replacement therapy, the researchers found no suggestion of a reduction in risk of heart attack: the odds ratio for the chance of heart attack among this group of women, as compared to nonusers of both NSAIDs and hormone replacement therapy, was 1.50.These findings suggest that hormone replacement therapy and NSAIDs might interact, with NSAIDs acting against a role for hormone replacement therapy in preventing heart attacks. At face value, these results are in conflict with the findings of one large trial, the WHI trial, which failed to find a benefit of HT in preventing heart attacks. However, a recent analysis of WHI suggests cardioprotective effects of HT in women close to the time of the menopause and this coincides with the younger age of women in the observational studies such as the present one rather than in the WHI overall. Observational research studies, such as the present one, are often difficult to interpret because the groups being compared are not necessarily equivalent. It's possible that women who take hormone replacement therapy, or NSAIDs, are in some way different from women who do not, which will bias the findings. Determination of the clinical implications of these findings would most appropriately be resolved in future trials, designed to address the question of interest.http://dx.doi.org/10.1371/journal.pmed.0040157.Please access these Web sites via the online version of this summary at menopausal hormone therapy, including links to information about the Women's Health Initiative trials, information about managing menopausal symptoms, and moreResources from the US National Institutes of Health on heart disease in womenResources from the US National Institutes of Health (MedlinePlus) about hormone replacement therapyInformation from NHS Direct, the UK National Health Service, about UK General Practice Research Database is the database utilized in this articleThe nonsteroidal anti-inflammatory drugs (NSAIDs) (note: Wikipedia is an internet encyclopedia anyone can edit)Wikipedia entry on  Indeed, deletion of the I prostanoid receptor undermined substantially the vascular benefit of estrogen therapy in ovariectomized mice rendered prone to atherogenesis by deletion of the low-density lipoprotein receptor [Premenopausal women are less susceptible to myocardial infarction (MI) and stroke than are males of the same age group, an advantage that is lost after menopause . Howeverreceptor .2 in vivo. Sufficient information is currently unavailable to address this question for selective inhibitors of COX-2. However, traditional (t) nonsteroidal anti-inflammatory drugs (NSAIDs), such as ibuprofen, variably inhibit both COX-1 and COX-2, unlike low-dose aspirin, which targets preferentially, albeit not exclusively, COX-1 [These observations prompted us to address the possibility that the failure to detect a benefit from estrogen might be partly attributable to a pharmacodynamic interaction with inhibitors of COX-2, the major source of PGIy, COX-1 ,5. The oThe General Practice Research Database contains computerized medical information entered systematically by general practitioners (GPs) in the UK and sent anonymously to the Medicines and Healthcare Products Regulatory Agency . The infn = 1,673) in the present study. We considered as fatal those individuals who died within the first 30 d after the occurrence of AMI and patients who died from CHD before reaching the hospital. Therefore in this study, designed a priori, AMI includes, if not otherwise stated, both fatal and nonfatal outcomes in the participants. A random date within the study period was generated for each of the study cohort members. All individuals with a random date included in their period of observation (from study entry to end of follow-up) were eligible as controls. A group of 20,000 control participants, frequency-matched by age, sex, and calendar year, was randomly sampled among with the study cohort. We applied to the controls the same computer-based exclusion criteria as we applied to AMI indviduals, using each participant's random date as his or her index date. For the present study, we included all women as control participants .We recently evaluated the association between tNSAIDs and acute MI , where m2, was calculated from the registered height and weight. Alcohol intake was used as directly registered by the GP. We also elicited the participants' use of the health services  in the two years before the index date.Information on coronary risk factors, comorbidities, and drug utilization was obtained from the database. Hypertension, diabetes, hyperlipidemia, smoking, cardiovascular disease, arthritis, and recent anemia were considered present when these specific diagnoses were registered in the database before the index date. History of CHD was defined by the presence of MI and/or angina. Cerebrovascular disease was defined by the presence of ischemic or hemorrhagic stroke. Body mass index, expressed in kg/mA list of all medications containing estrogens and/or progestogens recommended for HT and available in the UK during the study period was extracted from the British National Formulary. These drugs were grouped into the following regimens: (a) oral oestrogens; (b) transdermal oestradiol; (c) oestradiol implant; and (d) tibolone. In addition, oral oestrogens and transdermal oestradiol were classified as opposed  or unopposed  depending on whether a progestin was supplied along with oestrogens. Exposure to HT was classified into three groups: “current users” when the supply of the most recent prescription lasted until index date or ended in the year before the index date; “past users” when it ended more than one year before the index date; and “nonusers” when there was no recorded use before the index date. When current use was restricted to those exposed only in the month before the index date, the estimate for the interaction was slightly higher. There was considerable—73%—overlap in the women considered current users by the two definitions. The definition of current use for all other studied drugs, including tNSAIDs, was use of the drug in the month before the index date.A nested case-control analysis was performed to estimate the effect of HT as well as the interaction between HT and tNSAIDs on the risk of MI. We calculated the odds ratio (OR) and 95% confidence intervals (CIs) of MI using unconditional logistic regression. All estimates of risk were adjusted for age, smoking, hypertension, diabetes, obesity, hypercholesterolemia, ischemic heart disease, use of low-dose aspirin, and antihypertensive drugs.p = 0.0012. The figure also shows that these estimates of risk did not vary by duration of HT. The interaction was apparent in women taking both opposed and unopposed estrogens, with ORs of 1.39 (95% CI 0.65–2.95) and 1.86 (95% CI 0.79–4.42), respectively. Also, the interaction between HT and tNSAIDs was already evident within months of starting the tNSAID and was sustained, irrespective of the duration of tNSAID therapy or HT  . The corpy or HT . MenopauPrompted by these results, we sought the interaction in a secondary analysis of current and past HT users and in HT nonusers for patients on low- and high-dose aspirin, antihypertensives, antidepressants, and gastric acid–suppressing drugs—a further 15 comparisons. Clearly, it is possible that a significant result might be attained amongst such multiple comparisons by chance. Concomitant use of aspirin at 75mg/d, which preferentially targets COX-1 for inhibition, was associated with an OR of 0.71 (95% CI 0.26–1.94) approximating that in aspirin nonusers . However, the OR of a MI rose to 1.41 (95% CI 0.47–4.22) amongst users of HT with concomitant doses of aspirin 150 mg/d and above, where coincident inhibition of COX-2 should be more pronounced. We replicated these analyses adjusting for health services utilization, and the results were materially unchanged (unpublished data). We investigated whether this interaction was specific to tNSAIDs, and found that commonly used drugs, such as antihypertensive drugs, acid-suppressing drugs (proton pump inhibitors and/or histamine II antagonists), and tricyclic antidepressants failed to reveal an interaction with HT (unpublished data). Thus, the interaction between COX inhibitors and HT appeared to be specific. Also, when we performed a sensitivity analysis using a one-month time window instead of one year to define current use of HT, the OR of HT in women who were taking tNSAIDS concomitantly further rose to 1.85 (95% CI 1.01–3.41). Although we adjusted for matching factors  in our analysis and for a series of known cardiovascular risk factors, an unrecognized and therefore unmeasured factor that strongly related both to concomitant use of HT and NSAIDs on one hand, and to the risk of MI on the other might theoretically contribute to our results.The results of this study raise the possibility that concomitant medication with NSAIDs might undermine a cardioprotective effect of HT in perimenopausal women. Mechanistically, this might relate particularly to inhibition of COX-2. However, the present study pertained to tNSAIDs, which inhibit COX-2, but vary in their comcomitant inhibtion of COX-1 and did not include NSAIDs designed to attain specific inhibition of COX-2.Both the pattern of incidence and the manifestation of vascular disease differ between women and men. Premenopausal women exhibit a diminished burden of vascular disease compared with males of similar age; this difference is eclipsed after menopause . FurtherThe present study raises the possibility of a drug–drug interaction with cardiovascular implications involving COX inhibitors. Much controversy has surrounded the putative cardioprotective effects of HT reported in observational studies . Indeed,2 impacted particularly on initiation and early development of atherosclerosis [2 [Generally, when the results of observational studies are in conflict with an RCT, the balance of evidence favors the latter. While observational studies can control for recognized variables, they are vulnerable to an imbalanced distribution of unrecognized variables of relevance to the outcome of the study. On the other hand, WHI was also affected by some methodological constraints. The first was a differential detection bias . Close tclerosis . Thus, aclerosis —the apparosis [2 . Thus, w2. Scant information is presently available concerning the combined use of specific COX–2 inhibitors and HT. However, we examined a priori a database for an interaction between tNSAIDs and HT. Generally, the coincident, time-dependent inhibition of platelet COX-1–derived thromboxane A2 might be expected, if anything, to mitigate the impact of COX-2 inhibition by the tNSAIDs examined in this study [We have reported that estrogen dependent atheroprotection in mice depends in substantial part on COX-2 derived PGIis study . We do nis study ,23. Futuis study  is unsurIn summary, these observations, based on small numbers, are provocative rather than conclusive and are not intended to guide clinical practice, but rather to prompt additional research. However, they raise the possibility that coincident inhibition of COX-2 by tNSAIDs may undermine cardioprotective effects of HT in peri- and postmenopausal women. If evidence consistent with this interaction were obtained in other larger epidemiological datasets, the possibility could fruitfully be addressed directly in randomized controlled clinical trials.


In this ecologic study I examined ischemic heart disease (IHD) and diabetes mortality in rural agricultural counties of Minnesota, Montana, North Dakota, and South Dakota, in association with environmental exposure to chlorophenoxy herbicides, using wheat acreage as a surrogate exposure. I collected data on agricultural land use and 1979–1998 mortality from the U.S. Department of Agriculture and the Centers for Disease Control and Prevention websites, respectively. Counties were grouped based on percentage of land area dedicated to wheat farming. Poisson relative risks (RR) and 95% confidence intervals (CIs), comparing high- and medium- with low-wheat counties, were obtained for IHD, the subcategories acute myocardial infarction (AMI) and coronary atherosclerosis (CAS), and diabetes, adjusting for sex, age, mortality cohort, and poverty index. Mortality from IHD was modestly increased . Analyses of its two major forms were more revealing. Compared with low-wheat counties, mortality in high-wheat counties from AMI increased , and mortality from CAS decreased . Mortality from AMI was more pronounced for those < 65 years of age . Mortality from type 2 diabetes increased . These results suggest that the underlying cause of mortality from AMI and type 2 diabetes increased and the underlying cause of mortality from CAS decreased in counties where a large proportion of the land area is dedicated to spring and durum wheat farming. Firm conclusions on causal inference cannot be reached until more definitive studies have been conducted. Chemicaexposure . Resultsexposure , were siexposure . This siexposure  might alexposure .The chlorophenoxy herbicides 2,4-di-chlorophenoxyacetic acid  and 4-chloro-2-methylphenoxyacetic acid (MCPA) have been widely applied in the United States since World War II and are used for broadleaf weed control in wheat farming and maintenance of home lawns and parks, rights-of-way, and road sides . 2,4-D uResidents of agricultural counties of Minnesota, Montana, North Dakota, and South Dakota may be environmentally and/or occupationally exposed to 2,4-D and MCPA. Wheat acreage per county or percentage of the county’s land area dedicated to wheat farming was used in two previous ecologic studies as a surrogate measure of exposure to chlorophenoxy herbicides and/or contaminants, because information on herbicide use by county was not available , 2003. UIn the present ecologic study I investigated the possible links between environmental exposures to chlorophenoxy herbicides and mortality during 1979–1988 and 1989–1998 from IHD and type 2 diabetes. Counties with different levels of wheat farming provided a gradient of exposures, thereby overcoming the lack of a null referent. This hazard identification study can be one of the initial steps in defining an association between environmental exposures to chlorophenoxy herbicides and mortality from IHD and diabetes.International Classification of Diseases, 9th Revision (ICD-9) (I obtained data on underlying cause of death based on the  (ICD-9)  by state (ICD-9) . Only whFrom the combined 262 counties of the four states, agricultural counties with a mostly rural population were selected based on the criteria of ≥ 20% of county’s land area dedicated to cropland and ≥ 50% of population defined as rural U.S. . Rural pTo further define wheat acreage in rural, agricultural counties of Minnesota, Montana, North Dakota, and South Dakota as a surrogate for chlorophenoxy herbicide use, I determined acreage for the different classes of wheat in the selected counties. The percentage of herbicide-treated acreage for spring and durum wheat is larger than for winter wheat, which is a better competitor with weeds . For couI used Spearman correlations to determine the associations among 1964 and average 1970–1979, 1980–1989, and 1990–1999 levels of wheat acreage. Depending on the number of deaths available for the specific underlying cause under study, counties were grouped based on the median or tertiles of the percentage of land area dedicated to wheat. For diabetes mortality, I compared high-wheat counties with low-wheat counties. For mortality from IHD, including AMI and CAS, I compared high- and medium-wheat counties with low-wheat counties. For the univariate analyses by mortality cohort, sex, and age groups, I calculated standardized rate ratios (SRRs) and 95% confidence interval (CI) using direct age standardization based on the 1970 U.S. population, according to established methods . For thep < 0.0001), 0.95 (p < 0.0001), and 0.94 (p < 0.0001), respectively. This implied that a county with a high percentage of its land area dedicated to wheat during 1970–1979 was most likely also a high-wheat county during 1964, 1980–1989, and 1990–1999. The average total wheat acreage for 1970–1979 expressed as a percentage of a county’s land area was used as a surrogate measure of exposure to chlorophenoxy herbicides. Wheat acreage during the other time periods could probably also have been used given the high correlations. A USDA report showed that in 1976, > 90% of the herbicides applied to wheat in the United States consisted of chlorophenoxy herbicides  and 0.74 (67/91), respectively, before combining winter wheat acreage with durum and other spring wheat acreage by county to obtain a value for adjusted total wheat acreage.Characteristics of the 152 selected counties distributed over three groups based on tertiles of the counties’ wheat percentage are presented in The results from Poisson models adjusting simultaneously for mortality cohort, sex, age, and poverty are presented in I repeated the Poisson models after applying a correction factor based on the winter wheat acreage in Montana and South Dakota. The adjusted results were very similar to the nonadjusted results (data not shown).p-dioxins and dibenzofurans was observed in a German herbicide-producing plant . Long lag periods between time of exposure and time of diagnosis of the chronic disease under study may present problems . Use of The present ecologic study as well as several previous studies used existing databases . CompariIn summary, this ecologic study is an example of how population studies can make valuable contributions to public health by identifying potential environmental hazards.The population estimates used for the analyses were obtained from CDC WONDER and consist of Census intercensal estimates for 1971–1979 and 1981–1989, Census postcensal estimates for 1991–1999, and modified age–race–sex Census counts for the census years . These were the best population estimates available at the time of data collection (January–July 2003). Recent Census Bureau intercensal estimates for 1991–1999 showed that the postcensal estimates for 1991–1999 were too low. As a result, in the current study, the 1989–1998 population was underestimated by 0.6%. However, reanalysis based on the more accurate population data showed that this had negligible effects on the results (data not presented).


Our primary purpose was to determine the normal range and variability of blood volume (BV) in healthy children, in order to provide reference values during childhood and adolescence. Our secondary aim was to correlate these vascular volumes to body size parameters and pubertal stages, in order to determine the best normalisation parameter.Plasma volume (PV) and red cell volume (RCV) were measured and F-cell ratio was calculated in 77 children with idiopathic nephrotic syndrome in drug-free remission . BV was calculated as the sum of PV and RCV. Due to the dependence of these values on age, size and sex, all data were normalised for body size parameters.BV normalised for lean body mass (LBM) did not differ significantly by sex (p < 0.376) or pubertal stage (p < 0.180), in contrast to normalisation for the other anthropometric parameters. There was no significant difference between reference values for children and adults.LBM was the anthropometric index most closely correlated to vascular fluid volumes, independent of age, gender and pubertal stage. Alterations in blood volume (BV) lead to haemodynamic adaptations in acute and chronic diseases such as insulin-dependent diabetes mellitus, cardiac, renal and/or hepatic failure, and nephrotic syndrome -4. To evThe International Committee for Standardisation in Hematology has recommended measuring RCV by radioactive chromium or technetium methods, and PV by the radioiodine labelled albumin method . To dateThe aim of the present study was to obtain a reference frame for BV in children and adolescents, to calculate the best parameter for normalisation, and to evaluate the need for the use of the combined measurement method (i.e. PV plus RCV). We extended findings in our earlier cohort of patients , particuTwelve hours prior to administration of 131I-human serum albumin (HSA), KI was administered orally to prevent thyroid uptake of radioactive iodine. On the morning of the volume determination, 5 ml venous blood was drawn for red blood cell (RBC) labelling with 51 Cr , and theBV was calculated as RCV + PV , whole bIn males and prepubertal females (Tanner stage 1), LBM (a) = 0.407BW + 26.7H -19.2,In post pubertal or older females (Tanner stage 5), LBM (b) = 0.252BW + 47.3H - 48.3.In pubertal females (Tanner stages 2–4), LBM(c) = {LBM (a) + LBM (b)}/2These formulae were derived from the empirical relation between total body water (TBW), BW and H and between TBW and LBM (in kg) ,13. ElimThe study was approved by the ethical committee of University Hospital Ghent (1.5.643.98). All patients and/or their parents gave written informed consent.The study cohort consisted of 77 children with steroid responsive nephrotic syndrome, in whom vascular fluid volumes were measured during stable remission, and included the initial cohort of 31 children . All patData are expressed as mean ± standard deviation and compared using the Mann-Whitney U-test. Pearson's correlation coefficient and linear regression analysis among age, H, BW, BSA and LBM and the different vascular volume values  were used where indicated.2 (P70-P30) and mean BMI (w/h2) was 18.4 ± 4.2 kg/m2. Renal functional parameters, including glomerular filtration rate (GFR), renal plasma flow (RPF) and fractional excretion of sodium, were within the normal range [The population consisted of 42 males and 35 females; their baseline demographic and clinical characteristics are shown in Table al range . All patal range , indicat2 for BV normalised for LBM. The correlation coefficient for BV/LBM was also highly significant  in this paediatric population. The intercept of the regression line for BV normalised for LBM was close to 58 ml but did not pass through the origin , normalised for age, H, BMI, BW, BSA and LBM, are shown in Table in Table . Due to in Table . The regin Table . The indWhen we plotted the regression of BV as a function of LBM for the entire study population, in which the normal range was defined as twice the standard deviation and is identified by the dashed lines, we observed a continuum from puberty to adulthood Fig. . We obseTo our knowledge, this study is the largest to date on the direct determination of BV from PV and RCV in children and adolescents at various pubertal stages. Our pilot study in 31 patients aged 2–18 years, which measured PV and RCV, provided reference values for these parameters . The aimNa), and their plasma renin activity and aldosterone levels were within the normal range. In addition, long term follow up (> 6 y) showed no deterioration of renal function, thereby excluding severe renal parenchymal damage. This, together with the absence of clinical evidence for abnormal body composition, allows us to use this population as a reference for body fluid volumes in children. There is no evident reason why this highly selected study population cannot be considered as alternative and at least as the best reference frame available.These same ethical considerations are less applicable to nephrotic children in remission, since studies using radiolabelled compounds in these children have led to major changes in the diagnosis and treatment of hypo-/hypervolemia. Children with minimal change nephrotic syndrome in drug free remission are considered to have normal renal and haemodynamic parameters during proteinuria- and drug-free episodes ,15. GlomThe aim of this study was to determine a reference frame for body fluid volumes in children and to find the best normalisation parameters for BV. In the absence of very large study groups, normalisation is unavoidable, because of the large body size and age differences in children. The use of ratios as a consequence of normalisation, however, has pitfalls . In ordeWe were not surprised to observe a very good linear correlation between the different vascular volumes and the parameters related to size . Although correlations with other parameters (BW and BSA) were apparently limited, the larger intercept, as well as the sex and pubertal differences, make them unsuitable as simple normalisation parameters. Our finding of the highest linear correlation (R = 0.949) with LBM not only resolves the problem of sex differences, but is also in line with findings in adults . When plDespite statistical evidence in favour of the use of LBM, some critical comments must be made. Since the formula for LBM is derived from an empirical relationship, based on equations from a 20-year-old study, it is not known if this formula is still appropriate for children of this century. The equation is different for males/prepubertal females and pubertal/older females due to differences in body composition ,13,17, bBecause of these serious methodological considerations regarding the use of LBM, the other normalisation parameters should be evaluated as alternatives, although it is questionable whether children of all ages can or should be "forced" to fit a single normalisation. When we normalised for BSA, BW, or H, we observed differences between males and females during pubertal stages 2–4, as well as differences with prepubertal and adult values, suggesting that none of these alternatives can be used as a single normalisation parameter. Rather, we have to take into account sex and pubertal stage differences, even if we neglect the theoretical importance of a low intercept value. This is not surprising inasmuch as the concept of LBM was created to exclude gender differences in body composition, and, as a consequence, progressive pubertal changes. This is not just a mathematical trick, but also an unavoidable methodological necessity if we want to plot males and females in one reference frame. The only alternative would be to extend the study population to create reference frames dependent on age, sex and pubertal stage. The fact that this solution is not achievable due to many practical and ethical considerations favours the use of LBM as a single parameter.Statistical differences in the ratios of volume to various parameters of body composition may be related to differences in the derived body composition equations/nomograms rather than to true differences in RCV, PV, or BV. Significant differences in normalised values are apparently related to differences between the normalisation parameters during puberty, because BV was the same in every calculated ratio, and the best correlation was observed with derived parameters (LBM and BSA) rather than with measured variables . This is not surprising, because direct measured parameters are less suitable for normalising body composition and for evaluating pathophysiological mechanisms, including the adequacy of peritoneal dialysis, perspiration, and drug distribution volume.The aim of the present study was to determine a reference frame for blood volume studies in children, using a combination of RCV and PV measurements. The wide range of F-cell ratios clearly shows that the use of plasma volume and a fixed F-cell ratio of 0.91 is not feasible. Despite its major theoretical disadvantages, the use of adapted formulas for LBM emerged as the only statistically defensible normalisation parameter, due to its high correlation factor and low intercept. LBM also has the advantage of correcting for differences between the sexes and the pubertal stages, as well as being in accordance with findings in adults. The adaptation of the LBM equation during puberty, while artificial, may resolve a methodological problem. This study offers an important tool for studying conditions in which alterations in BV may be important, including hypertension, diabetic nephropathy, capillary leak syndrome, and nephrotic and nephritic syndromes.BSA: body surface areaLBM: lean body massBV: blood volumePV: plasma volumeRCV: red cell volumeTBW: total body waterBW: body weightP: percentileKI: potassiumiodideThe author(s) declare that they have no competing interests.AR contributed significantly in the design of the project, including formulation of questions and methods and design of the individual experiments, in patient recruitment, in performing the tests and experiments, in collecting the data, in the statistical analysis, in interpretation and discussion of the results and in writing the manuscript. SVA participated in the coordination of the study. MC participated in patient recruitment and in discussion of the results. RD contributed to collaboration in writing the manuscript, statistical analysis and correction of the manuscript. JVDW contributed significantly in the design of the project and its coordination, in statistical analysis, in presentation and interpretation and discussion of the results and in correction of the manuscript.The pre-publication history for this paper can be accessed here:


The aims with the present study are to analyze inter-population variations in serum concentrations of 2,2',4,4',5,5'-hexachlorobiphenyl (CB-153) and 1,1-dichloro-2,2-bis (In study populations of 3161 adults, comprising Greenlandic Inuits, Swedish fishermen and their wives, and inhabitants from Warsaw, Poland and Kharkiv, Ukraine, serum concentrations of CB-153 and p,p'-DDE, were analysed by gas chromatography-mass spectrometry.The median serum concentrations of CB-153 were for male and female Inuits 200 and 110, for Swedish fishermen 190 and their wives 84, for Kharkiv men and women 44 and 27, and for Warsaw men and women 17 and 11 ng/g lipids, respectively. The median serum concentrations of p,p'-DDE were for Kharkiv men and women 930 and 650, for male and female Inuits 560 and 300, for Warsaw men and women 530 and 380, and for Swedish fishermen 240 and their wives 140 ng/g lipids, respectively. The correlation coefficients between CB-153 and p,p'-DDE varied between 0.19 and 0.92, with the highest correlation among Inuits and the lowest among men from Warsaw. Men had averagely higher serum concentrations of CB-153 and p,p'-DDE, and there were positive associations between age and the POP-biomarkers, whereas the associations with BMI and smoking were inconsistent. Dietary seafood was of importance only in the Inuit and Swedish populations.CB-153 concentrations were much higher in Inuits and Swedish fishermen's populations than in the populations from Eastern Europe, whereas the pattern was different for p,p'-DDE showing highest concentrations in the Kharkiv population. The correlations between the POP-biomarkers varied considerably between the populations, underlining that exposure sources differ and that the choice of representative biomarkers of overall POP exposure has to be based on an analysis of the specific exposure situation for each population. Age and gender were consistent determinants of serum POPs; seafood was of importance only in the Inuit and Swedish populations. The use of persistent organochlorine pollutants (POP), such as polychlorinated biphenyls (PCB), polychlorinated dibenzo-p-dioxins, polychlorinated dibenzofurans and pesticides such as dichlorodiphenyl trichloroethane (DDT) has been restricted and bans implemented in many countries during the 1970's and 1980's, but e.g. DDT is still used in some areas for malaria vector control. These ubiquitous lipophilic POPs are highly resistant to both abiotic and biotic degradation, and are transported long distances by air and water streams.Some POPs can disrupt multiple endocrine pathways and induce a wide range of toxic responses . A varie) which aims to identify and characterize the impact of dietary POP on human fertility in an epidemiological setting [p-chlorophenyl)-ethylene  were in the same order of magnitude [The present study is a part of a collaborative research project funded by EU  with both total PCB concentration in plasma and serum from Swedish subjects and Inuits from Greenland -15. AnotThe aims with this publication are to a) analyze inter-population variations in CB-153 and p,p'-DDE exposure concentrations and b) assess inter-population variations in biomarker correlations, c) and finally to assess the relative impact of different determinants  for the inter-individual variations in serum concentrations of CB-153 and p,p'-DDE.The aim was to recruit 600 pregnant women and their male spouses in Greenland, Warsaw, and Kharkiv, respectively, and 200 Swedish fishermen and 600 Swedish fishermen's wives. A general criterion for eligibility was that the participants had to be born in the country of study and to be at least 18 years of age.Greenland pregnant Inuit women from 15 municipalities and 4 settlements from all regions within the country were contacted and informed about the project. In total 893 pregnant women were listed at the local midwives, but 235 did not fulfill the inclusion criteria, 32 could not be contacted and 38 did not want to participate. Blood samples were drawn from the pregnant women and their spouses , and interviews were made with both partners from June 2002 to December 2003. Of the 598 women included in the time to pregnancy (TTP) study (the results will be published elsewhere), 62 were living alone, 36 men could not be contacted, 34 men did not want to be interviewed, and 5 men did not want to give a blood sample. Moreover, for unknown reasons, blood samples were lost for 16 women and 22 men. Thus, in total, blood samples and interview data were available for 439 men (82 % participation rate) and 572 women (87 % participation rate)  women and 257 (37 % participation rate) men donated a blood sample and were interviewed from September 2002 to March 2003  women and 287 (12 % participation rate) men donated a blood sample and were interviewed from April 2003 to December 2004. The population had a mixed urban and rural background.Altogether 2478 pregnant women and their spouses who visited one of eight antenatal clinics or three maternity hospitals in Sweden was not focused on ongoing pregnancies. Instead blood samples were drawn from 189 professional fishermen from the Swedish east and west coasts cohorts that participated in a semen study between March 2001 and November 2001, and March 2002 and September 2002 , smoking habits, and intake of seafood. For the Inuits seafood comprised sea mammals as well as fish, whereas for the other populations it only refers to fish. For the fishermen's wives from Sweden information on life-long total length of breast-feeding was obtained and for the other women whether it was the first time they were given birth (primipara). For the populations from Sweden and Greenland we also used information regarding area of living. Information on current smoking was not obtained for the women from Sweden and information on current intake of seafood was not obtained for the Swedish fishermen.g for 15 min. Serum was transferred with ethanol rinsed Pasteur pipettes to ethanol rinsed brown glass bottles . A piece of aluminum foil was placed on top of the bottles which were then sealed. Sera were stored at -20°C until shipment, but it was accepted to keep it in refrigerator for up to four days.Blood samples were drawn from a cubital vein into 10 ml vacuum tubes for serum collection without additives . After cooling to room temperature the tubes were centrifuged at 4000 The sera were transported on dry ice to the Department of Occupational and Environmental Medicine in Lund, Sweden, where all analysis of CB-153 and p,p'-DDE were performed. The CB-153 and p,p'-DDE were extracted from serum by solid phase extraction using on-column degradation of the lipids and analysis by gas chromatography mass spectrometry as previously described . The relSerum concentrations of triglycerides and cholesterol were determined by enzymatic methods using reagents from Roche Diagnostics . The inter-assay CVs for cholesterol and triglyceride determinations were 1.5–2.0 %. The average molecular weights of triglycerides were assumed to be 807. For cholesterol we used an average molecular weight of 571, assuming that the proportion of free and esterified cholesterol in plasma was 1:2. Based on a paper by Rylander et al., the total lipid concentration in serum (g/L) was calculated by the following equations :Men: Total = 0.96 + 1.28*(triglycerides + cholesterol)Women: Total = 1.13 + 1.31*(triglycerides + cholesterol).Separate analyses for each population were performed for men and women. If the concentration was less than the limit of detection, the concentration was set to half of the detection limit based on the fresh weight concentrations. For CB-153 there were 122  and for p,p'-DDE there were 13  values below the limits of detection. The impact of potential determinants for inter-individual variations in CB-153 and p,p'-DDE serum concentrations were assessed by linear regression models. We also evaluated whether log transformations of the POP variables better fulfilled the model assumptions, which were checked by analysis of residuals according to the methods suggested by Altman . The det2 obtained from SPSS™), after successive addition of determinants to the final multivariate model, are given.The objective of the multivariate model building was to evaluate the impact of each of the determinants. In the first step we evaluated the associations for one determinant at a time with the CB-153 and p,p'-DDE concentrations in serum, respectively. In the second step we kept the determinant with the lowest p-value (if ≤ 0.15) and tentatively included all other determinants, one at a time. In the third step we kept the two determinants with the lowest p-values (if both ≤ 0.15) and tentatively included the remaining determinants, one at a time. This procedure was continued as long as the p-values for the all included determinants were ≤0.15. For the populations from Greenland and Sweden (only among the women) we also evaluated whether area of living modified the effect of seafood by including an interaction term in the models. The explained variances  and we did therefore not make age-adjusted comparisons for the women from Sweden.The correlations between serum concentrations of CB-153 and p,p'-DDE for the different populations were evaluated by Spearman's correlation coefficients.In total, the material included lipid adjusted serum concentrations of CB-153 and p,p'-DDE for 1172 men and 1989 women. In Tables For the men the mean serum concentrations of CB-153 was in the following order: Inuits (highest), Swedish fishermen, Warsaw and Kharkiv (lowest) , whom in turn had higher concentrations than the women from Warsaw .The women from Kharkiv had significantly higher age-adjusted serum concentrations of p,p'-DDE as compared with the Inuit women and the women from Warsaw . There was, however, no age-adjusted difference between the Inuit women and the women from Warsaw (p = 0.17).s) between serum concentrations of CB-153 and p,p'-DDE in the different populations varied between 0.19 and 0.92, with the highest correlation among Inuits and the lowest correlation among men from Poland  and Swedish fishermen (β = 9.0) as compared with men from Warsaw (β = 1.2) and Kharkiv (β = 2.0) (Tables There was a positive association between ) Tables , 8. The ) Tables , 8.BMI and smoking the pattern was very inconsistent, e.g. the effect of BMI on serum concentrations of p,p'-DDE went in different directions among the Swedish fishermen and the men from Warsaw  for the POP concentrations  as well as with the serum concentrations of p,p'-DDE (p < 0.001) (Table Among Inuit men the number of 1) Table . Among t1) Table . The intgiving birth to their first child had in general higher concentrations of POP concentrations, although most p-values were above 0.05  was one order of magnitude higher than the median CB-153 concentration (149 ng/g lipid) . The moss 0.92 and 0.93, respectively; Table s 0.79 and 0.76, respectively) support a common exposure source for both contaminants. This was not that obvious for the populations from Warsaw and Kharkiv, for which there were no previous data concerning the inter-correlation between CB-153 and p,p'-DDE. These large differences in correlation underline that the choice of representative biomarkers of overall POP exposure has to be based on an analysis of the specific exposure situation for each population. It might seem paradoxical that the correlation coefficients between the two exposure biomarkers became considerably lower when pooling the four populations, but the reason is that the exposure ranges did vary quite substantially between the different populations. Undoubtedly, it would have been valuable to have a more extensive exposure profile of the POPs than the two biomarkers analysed in this project, and we plan to perform such analyses on a stratified sub-sample from the study populations. On the other hand, it should be kept in mind that for practical and economical reasons it would have been impossible to analyse more than 3,000 samples, as in the present project, for a more detailed POP exposure profile. When performing epidemiological studies on reproductive outcomes, large enough sample sizes are needed to ensure a reasonable statistical power. Thus, a trade off is needed between costs for obtaining large enough number of samples and costs for detailed analyses of each sample, in order to obtain the most efficient study design.A noteworthy result of the present study was that the degree of correlation between the two POP biomarkers, CB-153 and p,p'-DDE, varied considerably between the different populations. The extremely high correlation coefficients  and the North Sea (west coast), respectively. For both male and female Inuits there were clear positive association between consumption of seafood and the POP markers. Among the fishermen's wives from Sweden the area of living modified the effect of fish consumption on POP concentrations. The impact of seafood  was greater among women from the Swedish east coast, which is explained by the higher concentration of POP contamination in fatty fish from the Baltic Sea as compared with fish from the North Sea at the west coast [When evaluating the impact of age on serum concentrations of POP in women, one has to keep in mind that age is associated with r's milk -27, as wr's milk  have prest coast . In a prst coast  and in Ust coast , fish cost coast . The seaBMI and serum concentrations of CB-153 or p,p'-DDE. The positive association between BMI and p,p'-DDE among the Swedish fishermen was the only significant finding, but the explained variance in the multivariate model did only increase with 1 % by including BMI in the model. In a previous study POPs were negatively associated with BMI among young and middle-aged male Inuits from Greenland [There was in the present study no consistent association between reenland , but we reenland ,30,31, breenland ,32. Two reenland ,31, wherreenland ,24. A poreenland . The assreenland . For somreenland . If thisreenland .smoking and serum concentrations of both CB-153 and p,p'-DDE. However, by including smoking in the multivariate model the explained variance did only increase with 1 %. In contrast, a negative association between current smoking and p,p'-DDE was observed among the women from Warsaw, whereas no association between POP markers and smoking was observed among women from Kharkiv. In Inuit men and men from Kharkiv, smoking was not associated with any of the POP biomarkers, whereas there was a negative association between smoking and CB-153 among Swedish fishermen. Thus, is it possible to explain this ambiguous pattern of associations? Our results are in accordance with previous studies of Inuit women from Greenland [Among the Inuit women we observed positive associations between current reenland ,34, and reenland . It is, reenland ,29. In creenland . It has reenland , but thiThe participation rate varied considerably between the different populations (10–87 %). We have, however, no reason to suspect that the subject's decision to participate was affected by any knowledge of their individual POP exposure. It was not possible to adjust for participation rate as this is a population characteristic and not an individual characteristic. Concerning the external validity of the present study, our conclusions are restricted to the age groups studied. Thus, we cannot say anything about serum concentrations of POP biomarkers in children, adolescents or in women past their reproductive age (except for Sweden). Three of the female study populations comprised pregnant women, which could hamper a comparison with non-pregnant women of the same age. However, gestational length affected the POP biomarker concentrations only in the Kharkiv population. Adjusting for the other determinants, each extra week of pregnancy in Kharkiv women was associated with a decrease of the POP biomarkers with 1 %. We think that the participating men from Greenland, Warsaw and Kharkiv were representative for the general population in the same geographical area, whereas the Swedish fishermen were more highly exposed to POPs than the general Swedish population.We used CB-153 and p,p'-DDE in serum as index biomarkers of exposure to POPs and there are several previous studies supporting this approach -15. It mPOP concentrations in blood have previously been analyzed in several studies both from Greenland and Sweden. Samples drawn in 1994–1997 from pregnant women from the Disco Bay area at the west coast of Greenland showed that both CB-153 and p,p'-DDE were at that time found in about 50 % higher concentrations than in the present study ,39. GlynSerum concentrations of CB-153 were much higher in the Inuit and Swedish fishermen's populations as compared with the populations from Eastern Europe, whereas the exposure pattern was different for p,p'-DDE showing highest concentrations in the Kharkiv population and relatively low concentrations in the Swedish fishermen's populations.The degree of correlation between the two POP biomarkers varied considerably for the different populations, which underlines that the exposure sources obviously differ between countries and that the choice of representative biomarkers of overall POP exposure has to be based on an analysis of the specific exposure situation for each population.Age and gender affected the serum POP concentrations in all populations whereas the associations with BMI and smoking were inconsistent. Dietary seafood and area of residence were of importance for the POP biomarker concentrations only in the Inuit and Swedish populations.BMI body mass indexCB-153 2,2',4,4',5,5'-hexachlorobiphenylCV coefficient of variationp-chlorophenyl)-ethylenep,p'-DDE 1,1-dichloro-2,2-bis (DDT dichlorodiphenyl trichloroethanePCBs polychlorinated biphenylsPOPs persistent organochlorine pollutantsTTP time to pregnancyThe author(s) declare that they have no competing interest.JPB, AG and LH initiated the project. JPB was main responsible for raising funding for the project. ARH, GT, HSP, JKL, KG, and VZ have been responsible for collecting all blood samples and for obtaining the interview data. BAGJ and CL were responsible for the chemical analyses of the POP biomarkers. GT had main responsibility for creating the joint database. LR performed the statistical analyses. LH, BAGJ, LR, JPB, GT and ARH helped to draft the manuscript. All authors participated in the design of the study and have read and approved the final manuscript.