A low ankle-brachial index (ABI) is associated with increased risk of coronary heart disease, stroke, and death. Regression model parameter estimates may be biased due to measurement error when the ABI is included as a predictor in regression models, but may be corrected if the reliability coefficient, R, is known. The R for the ABI computed from DINAMAP™ readings of the ankle and brachial SBP is not known.A total of 119 participants in both the Atherosclerosis Risk in Communities (ARIC) study and the NHLBI Family Heart Study (FHS) had repeat ABIs taken within 1 year, using a common protocol, automated oscillometric blood pressure measurement devices, and technician pool.The estimated reliability coefficient for the ankle systolic blood pressure (SBP) was 0.68  and for the brachial SBP was 0.74 . The reliability for the ABI based on single ankle and arm SBPs was 0.61  and the reliability of the ABI computed as the ratio of the average of two ankle SBPs to two arm SBPs was estimated from simulated data as 0.70.These reliability estimates may be used to obtain unbiased parameter estimates if the ABI is included in regression models. Our results suggest the need for repeated measures of the ABI in clinical practice, preferably within visits and also over time, before diagnosing peripheral artery disease and before making therapeutic decisions. The ratio of the ankle to the brachial systolic blood pressure (SBP), the ankle-brachial index (ABI), is an indicator of atherosclerotic vascular disease in the lower extremities and a simple, non-invasive measure of subclinical atherosclerosis -9. The AR, the reliability coefficient. Parameter estimates for additional covariates included in the model may be biased in any direction. Unbiased parameter estimates may be obtained using Stein estimators of true values [R is known [Regression parameters may be biased when predictors are measured with error ,35. The e values  or otheris known ,35.To measure ankle SBP, Doppler ultrasound devices have been used for many years to detect blood flow distal to an occluding cuff. Many epidemiologic CVD studies have included an ABI measured using either Doppler at the ankle and sphygmomanometry at the arm, or Doppler at both the ankle and the arm. In order to reduce interobserver variation, several studies including the Atherosclerosis Risk in Communities (ARIC) study  and the DINAMAP™ ankle and brachial SBPs were taken during ultrasound exams using the same protocol, equipment, and pool of sonographers in the ARIC study and the FHS. Using repeat measures of the ABI taken within a year apart for each of these studies, we estimate variance components and reliability coefficients for ankle and brachial SBPs and for the ABI. We also examined the effect of participant characteristics on the ABI reliability.The ARIC study cohort comprises 15,792 members aged 45–64 years of randomly selected households in four United States communities: Forsyth County, North Carolina; northwest suburbs of Minneapolis, Minnesota; Washington County, Maryland; and African American residents of Jackson, Mississippi ,39. ARICSeven hundred ten ARIC study participants and their families participated in the FHS clinic examination: 267 randomly-selected families, 340 families with high family risk scores, and 103 black participants at Forsyth . One memReplicate ankle-brachial index measures taken for each of the two studies were available for 335 ARIC probands, 120 of which were taken within 365 days. The ARIC ABI measure closest to the FHS exam was chosen; 48 participants had 3 measures, but in all cases the second and third measures were > 365 days apart. One participant who had a difference between ABI measures that was more than 3 standard deviations (SDs) from the mean pair difference was excluded. Of the remaining 119 participants, 25 (21 percent) had their first measure in the FHS and the second at ARIC visit 3 and 22 (18 percent) had the second at ARIC visit 4; 72 (61 percent) had the first ABI measure at ARIC visit 3 and the second at the FHS exam. Measures were taken an average of 228 days apart.Ankle and brachial blood pressures were obtained immediately prior to the ultrasound examinations by the same pool of 11 sonographers in both the FHS and ARIC follow-up visit clinic exams, using the same protocol and equipment ,41. WithAt the ARIC baseline survey, the ankle SBP was measured with the participant prone before and after ultrasound scanning of the popliteal artery, then brachial SBPs were measured supine every 5 minutes during the carotid ultrasound. The ABI was computed as the ratio of the average of the last two available ankle SBPs to the average of the first two brachial SBPs.In the ARIC study and the FHS, hypertension was defined as a sitting SBP > 140 mmHg, diastolic blood pressure > 90 mmHg, or use of antihypertensive medication within 2 weeks prior to the examination. Diabetes was defined as a fasting glucose level ≥ 7.8 mmol/L (140 mg/dL), a non-fasting level ≥ 11.1 mmol/L (200 mg/dL), self-reported history of diabetes, or the use of hypoglycemic agents. In ARIC, prevalent CHD was defined as evidence of a prior myocardial infarction on a 12-lead ECG, or self-reported history of a physician-diagnosed heart attack, coronary bypass surgery, or coronary angioplasty. In the FHS self-reported, physician diagnosed coronary heart disease was confirmed by review of hospital records.p values < 0.05 were considered statistically significant. Characteristics of the ABI reliability study subjects were compared to characteristics of the ARIC source population at visits 1 and 3 using Kruskal-Wallis two-sample tests for continuous variables and Wilcoxon tests for categorical variables.For all statistical tests, two-sided R, was computed as the ratio of the between-person variance to the total variance and can be interpreted as the proportion of the total variance not due to within-person variation, or as the correlation between measures made at repeat visits for an individual. The within-person coefficient of variation (CV) was computed as 100 times the square root of the within-person variance divided by the mean of the individual mean replicates. These methods were applied to ankle and brachial SBP measurements as well as to ABI. In addition, the multivariate generalization of this method was applied to produce the multivariate error covariance matrices and between-person covariance matrices for ankle and arm SBP and the component-wise ratios of the error and between-person covariance to total covariance. .Each of the two observed values was assumed to represent the sum of the true value and random within-person variation . Within-Because the arm SBPs and ABIs were not normally distributed, confidence intervals for the estimates of variance components and reliability coefficients were derived from 300 bootstrap samples . We compTo examine the assumption of independence of the within-person variation from the ABI level, we compared the estimated square root of the within-person variance  for those whose first of the two repeat ABIs was at or below with those whose first ABI was above the median ABI in the 119 reliability study subjects. We also examined plots of the pair differences versus the pair means for the ankle SBP, arm SBP, and ABI .The ABI measured at the ARIC baseline survey is of interest as a predictor of subsequent events. Because repeat ankle and arm SBPs are available for the baseline survey, the ABI is usually computed as the ratio of the average of two ankle SBPs (in the same ankle) to the average of two arm SBPs (in the right arm). We estimated the within-visit multivariate error and total covariance matrices using repeat ankle and arm SBP measures taken at the ARIC baseline survey. The ratios of within-visit multivariate error to total covariance matrices from the baseline survey data were multiplied times the total variance in the ARIC/FHS repeat data to partition the non-between person variance found there into between-visit and within-visit components. We simulated a two-measures-per-visit repeatability dataset very much like the actual repeatability dataset by taking each subject's mean of the two actual repeated BP measures as the "true" BP values, and adding independent random "between-visit" and "within-visit" Gaussian measurement error, of size to make these two variance components the same proportion of the total variance as in the results of the repeatability study and the baseline survey within-visit repeatability analysis. With the simulated data, the reliability coefficient for the ABI was computed as a ratio of means, along with the reliability coefficient using just the first measure at each visit. This was repeated 1000 times and the reliability coefficients averaged.2, and mean ABI 1.16; 13 (10.9 percent) had diabetes mellitus, 43 (36.1 percent) had hypertension, and 31 (26.0 percent) had coronary heart disease. Characteristics of this group were similar to those of the ARIC cohort in Minneapolis and Forsyth County at the time of the ARIC baseline survey and at visit 3, except with respect to the distribution of characteristics related to the FHS sampling procedures: race/ethnicity, field center, and CHD prevalence  were women, 100 (84.0 percent) were white, and 77 (64.7 percent) were from the Forsyth County field center. At the time of the first ABI measure, the mean age was 61.3 years, mean weight 75.7 kg, mean body mass index 28.0 kg/mThe estimated reliability coefficient for the ankle SBP was 0.682 : 0.570, 0.772), for the arm SBP was 0.736 , and for the ABI was 0.612  (Table p = 0.11), which resulted in a statistically significant average increase of 0.031 in the ABI (p = 0.02). Adjustment for the fixed effect of time (first v. second measure) in an analysis of variance model changed the variance component estimates very little. FHS ankle SBP measures were an average of 3.2 mmHg higher than ARIC measures (p = 0.10). ABI measures in the FHS were 0.029 higher, on average, than ARIC measures (p = 0.03); controlling for the fixed effect of time reduced the study difference to 0.024 (p = 0.09). No significant effects of time or study were apparent for arm SBP.The ankle SBP increased an average of 3.1 mmHg between the first and second measures (2 than in those with a BMI > 27 kg/m2 (0.083 versus 0.126), resulting in a significantly greater reliability coefficient in those with a lower compared to those with a higher BMI (0.741 versus 0.495). Those whose same ankles were measured had a statistically significantly lower reliability coefficient (R = 0.486) than those with different ankles measured (R = 0.706). There were too few African American subjects, subjects with diabetes, and subjects who initiated or discontinued hypertension treatment to test for subgroup differences. Within-person SDs and reliability coefficients did not differ significantly by age ≤ or > 62 years, gender, hypertension, or by whether the time between measures was greater or less than the median interval of 240 days. Except for the estimated total variance, estimates did not differ significantly between those in the high family risk score and the random sample FHS sampling groups; the proportion of reliability study subjects in the African American sampling group was too small to test for differences.The estimated ABI variance components differed in some subgroups Table . The witWithin-person SD estimates did not differ significantly between those with an ABI ≤ 1.15 and those with an ABI > 1.15 (0.113 versus 0.097), the median ABI in the 119 reliability study subjects. A plot of the pair differences versus the pair means revealed no obvious dependence of the spread of the differences on the ABI level Figures , 3; howeThe multivariate within-person error covariance and true-to-total ratio covariance matrices for ankle and arm SBPs estimated from the reliability study are given in Table The multivariate within-person (between-visit) error covariance and true-to-total component-wise ratio covariance matrices estimated from the ARIC baseline survey are given in Table From ARIC baseline survey data, we also estimated the within-visit variance and reliability of repeat ABIs, arbitrarily computed as the ratio of the first ankle to first arm SBP and the ratio of the second ankle to second arm SBP. The within-visit error variance for the ABI was 0.00636, and the reliability coefficient was 0.681.The ABI measured at the ARIC baseline survey, computed as the ratio of the average of two ankle SBPs to the average of two arm SBPs, is of interest as a predictor in regression models. Using the estimates of the total and within-visit errors , but is lower than that found for other measures. A reliability coefficient of 0.60 was estimated for the mean carotid IMT, averaged over 3 arterial sites bilaterally, in 279 subjects who participated in both the ARIC study and the FHS and with carotid B-mode ultrasounds within a year apart . A reliability coefficient of 0.67 was estimated for the mean carotid IMT measured at three visits, 7–14 days apart in 36 volunteers from each of the four ARIC field centers ; measureThe within-person variation in this reliability study includes variation due to within and between observer, day, and leg, and physiologic variation. Thus, the ABI reliability estimated from the reliability study data is lower than the within-visit reliability estimated from ARIC baseline survey data, because the reliability study's within-person variability includes both between-visit and within-visit variation. The between-visit variation includes variation due to different technicians, different machines, and within-person biologic variation.The reliability of the ABI was lower than that of either of its components: the ankle SBP or arm SBP. The lower reliability of the ankle SBP compared with that of the arm SBP may be due to the greater difficulty in placing a cuff on the more conical extremity.Estimates of within-person variation in this reliability study apply to measures using the methodology at ARIC visits 3 and 4 and in the FHS, and may not apply to other situations. Extreme care was taken in these studies to minimize measurement variation, including a detailed, common protocol and training and certification of technicians. Reliability is likely to be lower in patient care settings, where these methods are unlikely to be standardized. Different oscillometric devices may vary in the algorithms employed for determining systolic blood pressure  and the The reliability study defined the ABI as a ratio of single blood pressure measurements. In the ARIC baseline survey, the ABI is defined using the mean of two ankle SBPs (in the same leg) and the mean of two arm SBPs (in the same arm), where the measures were taken approximately 5 minutes apart, and in the study of associations between the ABI and potential disease outcomes this ABI will be used. We have seen that the reliability coefficient for such an ABI is notably higher than that computed directly from the ARIC/FHS reliability study.n measures, this interval becomes the ABI average ± 0.21/√n, so the interval for an ABI based on the average of two visits would be ABI ± 0.15, and average ABI ± 0.12 if based on 3 measures. The reliability of oscillometric measurement of the brachial SBP is not much greater than that of the ABI, and multiple measures of the brachial SBP are recommended before instituting hypertension treatment [In clinical practice, repeated measures of the ABI should be made, preferably within visits and over time before diagnosing peripheral artery disease and before making therapeutic decisions. Based on the within-person variance estimated here, the 95% confidence interval (1.96 times the within-person standard deviation) for one person based on one measure of the ABI would be the ABI ± 0.21. In general, for the average of reatment .Most population-based epidemiologic studies have measured the ankle pressure in both legs, once  or twicep < 0.001); Forsyth also had a higher proportion of subjects with diabetes (14 percent v. 5 percent) than Minneapolis. The center difference might also reflect different proportions of subjects with higher within-person variation, although the difference in reliability coefficients between the two centers was not larger than chance alone would explain and small numbers in these subgroups prevent further analysis. Counter to intuition, the ABI had a lower reliability in those whose ABI was measured in the same leg than in those whose was measured in different legs; this may represent a chance finding in this small study. It would have been desirable to examine whether the reliability differed in subjects with and without ABI-defined LEAD , but the limited sample size did not allow for this analysis.The limited number of reliability study subjects in certain subgroups limited our ability to test for subgroup differences. The significantly higher within-person SD at Forsyth might be because 7 technicians were employed at the Forsyth County field Center and 4 at Minneapolis. The Forsyth County field center had 25 percent African American subjects, while Minneapolis subjects were all white  declare that they have no competing interests.GH and BW conceived of and designed the study. BW and LC performed the statistical analyses. GH, LC, and BW interpreted the results. BW drafted the manuscript. All authors revised the manuscript for intellectual content, and read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


Questions regarding the distribution of stress in the proximal human femur have never been adequately resolved. Traditionally, by considering the femur in isolation, it has been believed that the effect of body weight on the projecting neck and head places the superior aspect of the neck in tension. A minority view has proposed that this region is in compression because of muscular forces pulling the femur into the pelvis. Little has been done to study stress distributions in the proximal femur. We hypothesise that under physiological loading the majority of the proximal femur is in compression and that the internal trabecular structure functions as an arch, transferring compressive stresses to the femoral shaft.To demonstrate the principle, we have developed a 2D finite element model of the femur in which body weight, a representation of the pelvis, and ligamentous forces were included. The regions of higher trabecular bone density in the proximal femur  were assigned a higher modulus than the surrounding trabecular bone. Two-legged and one-legged stances, the latter including an abductor force, were investigated.The inclusion of ligamentous forces in two-legged stance generated compressive stresses in the proximal femur. The increased modulus in areas of greater structural density focuses the stresses through the arch-like internal structure. Including an abductor muscle force in simulated one-legged stance also produced compression, but with a different distribution.This 2D model shows, in principle, that including ligamentous and muscular forces has the effect of generating compressive stresses across most of the proximal femur. The arch-like trabecular structure transmits the compressive loads to the shaft. The greater strength of bone in compression than in tension is then used to advantage. These results support the hypothesis presented. If correct, a better understanding of the stress distribution in the proximal femur may lead to improvements in prosthetic devices and an appreciation of the effects of various surgical procedures affecting load transmission across the hip. Despite recent advances in modelling, and the success of total hip replacements, it is still not clear what stresses are generated within the proximal femur (the head and neck) during physiological loading. The traditional description is based on the work of the nineteenth century engineer, Cullman, who observed the drawings of the anatomist, Meyer, and likened them to the stress pattern of a crane which he was currently analysing . He propHowever, a small number of dissenters have drawn attention to the number of muscles that cross the articulation of the hip. These muscles apply forces that pull the femoral head into the acetabulum and should not be ignored -6. In adMany subsequent studies have investigated the stress distribution within the human femur ,3,12-16,Given that bone is stronger in compression than tension ,22 and tAn FE model was constructed comprising the shaft and proximal femur articulating freely with a representation of the acetabulum Figure . The ace-1 [in vivo, but are the strains required in order that the link elements apply forces that would be expected to include those experienced in vivo. Because the femoral head was allowed rotational freedom inside the acetabulum, the position of the femur could be adjusted automatically during the solving process until equilibrium, and a solution, was achieved. The final ligamentous forces were always less than the failure load of the capsule reported in the literature [The FE model was developed using ANSYS 8.0 software . The model consisted of 2162 eight-node quadrilateral elements (PLANE82)  (476 N) in two-legged stance with a link element strain of 5%. This compares with between 0.59–1.0 BW for two-legged stance measured by Bergmann et al. , and eitInvestigating the one-legged stance analysed by Pauwels  suggesteThe hip, despite its apparent simplicity, is surrounded by a very complicated arrangement of muscles and ligaments. The joint capsule and its ligaments are recognized as being among the thickest and strongest ligaments of the body and they are always in tension to prevent dislocation of the hip; a rare event ,8. They The results of our model demonstrate that when ligamentous and muscular forces are included, the stresses in proximal femur are predominantly compressive. This would appear to make better use of the mechanical properties of bone, which is stronger in compression than in tension . CombiniThe author(s) declare that they have no competing interests.KER performed the FE analysis, participated in the design and planning of the project and wrote the first draft of the manuscript. JMR participated in the design and planning of the project, supplied the detailed technical expertise in FE modelling and helped to draft and revise the manuscript. RMA conceived the study, participated in its design and coordination and helped to draft and revise the manuscript. All authors read and approved the final manuscript.


To compare single- with two- layer intestinal anastomosis after intestinal resection: a meta-analysis of randomized controlled trials.Randomized controlled trials comparing single- with two-layer intestinal anastomosis were identified using a systematic search of Medline, Embase and the Cochrane Library Databases covering articles published from 1966 to 2004. Outcome of primary interest was postoperative leak. A risk ratio for trial outcomes and weighted pooled estimates for data were calculated. A fixed-effect model weighted using Mantel-Haenszel methods and a random-effect model using DerSimonian-Laird methods were employed.2 = 10.5, d.f. = 5, p = 0.06).Six trials were analyzed, comprising 670 participants . Data on leaks were available from all included studies. Combined risk ratio using DerSimonian-Laird methods was 0.91 (95% CI = 0.49 to 1.69), and indicated no significant difference. Inter-study heterogeneity was significant (χNo evidence was found that two-layer intestinal anastomosis leads to fewer post-operative leaks than single layer. Considering duration of the anastomosis procedure and medical expenses, single-layer intestinal anastomosis appears to represent the optimal choice for most surgical situations. The basic principles of intestinal suture were established more than 100 years ago by Travers, Lembert and Halsted ,3, but cHistorically, two-layer anastomosis using interrupted silk sutures for an outer inverted seromuscular layer and a running absorbable suture for a transmural inner layer has been standard for most surgical situations. Some recent reports have described single-layer continuous anastomosis using monofilament sutures as requiring less time and cost than any other method, without incurring any added risk of leakage -8. Many While numerous RCTs have addressed this issue, no published reports have described meta-analysis of RCTs to date -14. We tA search of any language literature was performed through August 2004 to identify reports of randomized controlled trials comparing single- with two-layer intestinal anastomosis. Data bases searched were Medline (1966-April 2004), the Cochrane Register of Controlled Trials , and EMBASE(1986-August 2004). A detailed description of the search strategy is provided in the Appendix see . ElectroThis meta-analysis included studies clearly describing the following: 1) study design ; 2) main outcome ; 3) target population ; and 4) availability of leak data. Studies that were not RCTs were excluded from the analysis.Each investigator decided independently which reports should be included for analysis. Any disagreement was settled by consensus between all investigators. Data were extracted independently by two investigators (SS and YT), with any disagreements resolved by a final reviewer (YN).Outcome of primary interest was risk of leakage related to intestinal anastomosis. Secondary outcomes comprised mortality, duration of anastomosis procedure, duration of total parenteral nutrition (TPN), length of hospital stay, risk of wound infection, and cost of sutures.Quality of primary studies was evaluated as described by Jadad et al . This me2 test (Q statistics)[The fixed-effect model weighted using Mantel-Haenszel methods was used for pooling the risk ratio , followeatistics). A homogatistics).Meta-regression analyses were performed to explore sources of heterogeneity. Variables comprising year of publication, mean age of study participants, and percentage of male patients were examined for significant effects on risk of leak. In addition to that, informal graphical exploration using a L'Abbe plot was made and sensitivity analysis was performed ,20. And The potential for publication bias was examined using the funnel plot method , and theAll statistical analyses were performed using STATA statistical software version 8.1 . We used a user-written add-on Stata routine "metan", which was written by Bradburn et al.. ResultsFigure The 6 trials included for analysis comprised 670 participants . Table Mean Jadad score was 1.7  .Data on leaks were available from all included studies. In the study by Goligher et al., almost half of the cases that had been allocated to the single-layer group experienced leakage (31 of 69 patients). Included studies showed a tendency toward total risk of leaks decreasing annually Table . Inter-sFor other outcomes, no meta-analysis could be performed due to lack of sufficient data. Data on mortality related to intestinal anastomosis, reported only in the study by Irvin et al, was equal in both groups. The arithmetical mean duration of anastomosis procedure in two included studies was 23.4 min vs. 36.9 min (single vs. two-layer), and arithmetical mean length of hospital stay was 9.9 days vs. 13.0 days. Data on duration of TPN, reported only in the study by Maurya et al., showed that the single-layer group could tolerate oral fluids earlier, and duration of intravenous alimentation was shorter compared to two-layer patients . No study reported risk of wound infection. Suture cost was reported only in the study by Burch et al., and was $4.5 for single-layer compared to $35.4 for two-layer anastomosis.2 = 2.96, d.f. = 4, p = 0.57). Another sensitivity analysis was performed including only high-quality studies (n = 3) with a Jadad score ≥2. The combined risk ratio for leak was 1.65 (95% CI = 1.04 to 2.61) with DerSimonian-Laird estimate, indicating more favorable results for two-layer anastomosis compared to single. In two included studies by Everett et al. and Ordorica et al., operations were performed only by consulting surgeons, staff surgeons, and residents of ≥5 years experience. Combined risk ratio for leak was 0.61 (95% CI = 0.28 to 1.35), and no significant difference was found.Meta-regression revealed that neither year of publication, mean age of study participants, nor percentage of male patients were related to risk of leak. From the graphical exploration using a L'Abbe plot Fig. , we consFunnel-plot, Begg's test and Egger's test were performed to evaluate potential for publication bias in terms of leakage rate related to intestinal anastomosis. The funnel-plot showed a symmetrical pattern, and neither statistical test revealed the presence of significant publication bias .The present study assessed the efficacy and safety of single- and two-layer anastomosis after intestinal resection. The main finding of the study was that there is no evidence of a difference in terms of risk of leak but that there is insufficient evidence to rule out a modest but potentially important difference. Sensitivity analysis excluding the study by Goligher et al. suggested it as the source of heterogeneity. In their trial, techniques of vertical mattress sutures in the posterior two-thirds of the circumferences and Lembert sutures of horizontal mattress type in the anterior third of the bowel circumference were performed in single-layer group and reported the highest risk of leaks (45%). One possible explanation of this high rate of leaks may be their inclusion criteria, high and low colorectal anastomosis. On this subject, they described "We are quite unable to explain the difference between Everett's results and ours" in their report . This suNone of the studies except Ordorica et al. met the requirements for appropriateness of double-blinding. In the study by Ordorica et al., neither the physician performing the assessments nor the pediatric patient knew the type of anastomosis. However, assessing outcomes under blinding is virtually impossible in surgical trials. We therefore regarded studies with a Jadad score of 3 as high-quality studies.There are several limitations in current study. First, the study by Goligher et al. had a substantial influence on the combined risk ratio. However, the main conclusion of a lack of evidence for an advantage of two-layer over single-layer anastomosis is unaffected, as the result of a sensitivity analysis excluding this study was more favourable to single-layer. Secondly, the quality of individual RCTs included in our analysis was not necessarily high, mainly due to a lack of blinding. This is, however, inevitable in most surgical trials. Thirdly, there were differences in inclusion criteria, definition of the term "leak" and suture techniques for studies included for meta-analysis. Exclusion criteria among studies also varied. Lastly, the total number of patients included for this meta-analysis might not have been sufficient to identify small differences between the two techniques. However, no significant differences in methods were identified. Future RCTs may yield different conclusions from meta-analysis. Despite these limitations, we believe that this meta-analytic overview provides the current best information in making clinical decision to choose a surgical suture techniques.The current meta-analysis clarified that two-layer intestinal anastomosis offers no definite advantage over single-layer anastomosis in terms of postoperative leak. Considering duration of the anastomosis procedure and medical expenses, single-layer intestinal anastomosis may prove the optimal choice in most surgical situations.RCTs = randomized controlled trialsTPN = total parenteral nutritionCI = confidence intervalThe author(s) declare that they have no competing interests.SS conceived the study, participated in the design of the study, searched the reports, selected and evaluated reports, performed data abstraction and drafted the manuscript. HY and TS participated in the design of the study, searched the reports, selected and evaluated reports. YT participated in the design of the study and performed the data abstraction and statistical analysis. YN participated in its design and coordination and helped to draft the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:A detailed description of the search strategy is provided in this file.Click here for file


To comply with regulatory policies and public health authorities recommendations, trans fats should be replaced in food products. The study by Sundram et al.  reporting the effect on CVD risk factors of interesterified fat (IE) and partially hydrogenated soybean oil (PHSO) compared to palm olein (POL) has been critically analyzed. The study design and in particular the composition of the tested fats was not suitable to properly answer the question raised regarding the effect of alternative ingredients to trans fats on plasma lipids. The observed effects are divergent with predicted data derived from the literature model consolidated using the individual results of 60 randomized clinical trials. The results of the study published by Sundram and co-workers have to be considered with awareness.Consumption of  Dear Editor,trans fatty acids (TFA) has been shown to be associated with increased risk of cardiovascular disease (CVD) . Reversed-phase high-performance liquid chromatography (RP-HPLC) analysis of TAG cannot discriminate TAG regioisomers. The difference between PPO and POP is that the oleic acid is esterified in Sn-1(3) or Sn-2 position on the Sn-glycerol, respectively. These structural data could be assessed by either partial deacylation of TAG using pancreatic lipase or Grignard reagent [Sundram and co-workers  hypothes reagent , by mass reagent  or by NM reagent .trans fats in food products are available. However, the study published by Sundram and co-workers does not allow concluding that palm oil is healthier than interesterified fats.The neutral effect of stearic acid, compared to oleic acid, on CVD risk factors such as HDL-C, LDL-C and diagnostic ratio has been recently confirmed by Thijssen and Mensink . These rCardiovascular disease (CVD); Confidence intervals (CI); High-density lipoprotein cholesterol (HDL-C); Interesterified fats (IE); Low-density lipoprotein cholesterol (LDL-C); Nuclear magnetic spectroscopy (NMR); Palm olein (POL); Partially hydrogenated soybean oil (PHSO); Triacylglycerols (TAG)The authors declare that they have no competing interests.Frédéric Destaillats, Julie Moulin and Jean-Baptiste Bezelgues participated equally to the critical analysis of the study published by Sundram and co-workers (3).


The accurate determination of exposure to environmental tobacco smoke is notoriously difficult. There have been to date two approaches to determining this exposure in the study of association of passive smoking and lung cancer: the biochemical approach, using cotinine in the main as a marker, and the epidemiological approach. Typically results of the former have yielded much lower relative risk than the latter, and have tended to be ignored in favour of the latter, although there has been considerable debate as to the logical basis for this. We settle this question by showing that, using the epidemiologically based meta-analysis technique of Wald et al. (1986), and misclassification models in the EPA Draft Review (1990), one arrives using all current studies at a result which is virtually identical with the biochemically-based conclusions of Darby and Pike (1988) or Repace and Lowry (1990). The conduct of this meta-analysis itself raises a number of important methodological questions, including the validity of inclusion of studies, the use of estimates adjusted for covariates, and the statistical significance of estimates based on meta-analysis of the epidemiological data. The best estimate of relative risk from spousal smoking is shown to be approximately 1.05-1.10, based on either of these approaches; but it is suggested that considerable extra work is needed to establish whether this is significantly raised.


We very much welcome the National Children’s Study, which promises to raise the study of factors affecting reproduction and development to a new level. An impressive and exciting range of new methodologies is being developed .However, we think it important to correct some of the inaccurate statements concerning the use of retrospective time to pregnancy (TTP) made by Retrospective studies are not necessarily pregnancy based. They can be conducted in random population-based samples and frequently are cross-sectional or birth cohort studies , therebyA central issue is planning bias, the question being how to exclude accidental (unplanned) pregnancies without bias occurring if the exposure variable is associated with the degree of “plannedness.” Retrospective studies can readily investigate this by following the standard guidance to collect full information for all pregnancies, including all covariates, and carry out parallel analyses with “unplanned pregnancy rate” as outcome variable . ProspecWe agree that a major limitation of retrospective studies is that it is impossible to obtain detailed, timed information on exposures and key biologic events such as ovulation, and difficult to ascertain certain covariates such as frequency or timing of intercourse. This is the key strength of the prospective design. On the other hand, retrospective studies are representative because, as already noted, sampling from the general population is available and planning bias can be handled. The questions are easily administered and answered, and the response rate is high. Even response bias can be avoided by nesting the TTP questions within a more general population survey, thus decoupling survey nonresponse from differential fertility or other motivation that would convert low response rates to response bias . SelectiNot only are prospective studies time-consuming and costly, and therefore likely to be rarely used, but they have important methodologic drawbacks. For example, it is impossible to distinguish the approximately 3% of couples who are sterile from those who merely take a long time to conceive , unless follow-up is extremely long.More seriously, prospective studies are dominated by the lack of a sampling frame  and by a potent combination of planning bias and response bias. They can include only couples who deliberately plan and are willing to volunteer for onerous monitoring. This is acceptable for internal comparisons  but raises serious problems with external validity.


In order to establish a consistent method for brachial artery reactivity assessment, we analyzed commonly used approaches to the test and their effects on the magnitude and time-course of flow mediated dilation (FMD), and on test variability and repeatability. As a popular and noninvasive assessment of endothelial function, several different approaches have been employed to measure brachial artery reactivity with B-mode ultrasound. Despite some efforts, there remains a lack of defined normal values and large variability in measurement technique.Twenty-six healthy volunteers underwent repeated brachial artery diameter measurements by B-mode ultrasound. Following baseline diameter recordings we assessed endothelium-dependent flow mediated dilation by inflating a blood pressure cuff either on the upper arm  or on the forearm .p = 0.0001). Measurement of the test repeatability showed that differences between the repeated measures were greater on average when the measurements were done using the proximal method as compared to the distal method .Thirty-seven measures were performed using proximal occlusion and 25 with distal occlusion. Following proximal occlusion relative to distal occlusion, FMD was larger  and elongated (107.2 s vs. 67.8 s, These findings suggest that forearm compression holds statistical advantages over upper arm compression. Added to documented physiological and practical reasons, we propose that future studies should use forearm compression in the assessment of endothelial function. Impaired endothelial function is recognized as an early and modulating process in the pathophysiology of atherosclerotic cardiovascular disease. EndotheTwenty-six volunteers participated in the study. Participants were age 18–49, healthy, normotensive, nonsmokers. The study was approved by the University of Washington Human Subjects Review Committee; subjects provided informed consent for all procedures.In the first 19 participants, FMD was measured twice on two separate days, at least two weeks apart. In the remaining seven participants, FMD was measured on two separate days, on two occasions on each day separated by one hour for all individuals.All measurements of brachial artery diameter and FMD were performed in the morning, in a quiet and dark room and at controlled ambient temperatures between 20°C and 26°C. The electrocardiogram (ECG) was continuously monitored. Female participants were examined during the follicular phase of their menstrual cycle. Studies were conducted after an overnight fast of at least 10 hours (water was permitted), with the subjects supine and after 10 minutes of rest. The subject's right arm was comfortably immobilized in the extending position, allowing for ultrasound scanning of the brachial artery 5–10 cm above the antecubital fossa.In each examination, recording of vessel images were followed by inflation of a cuff to suprasystolic pressure (40 to 50 mmHg above systolic pressure) for 5 minutes as suggested by findings in Corretti et al. Then thThe brachial artery was scanned in the longitudinal section using an HDI 5000 Ultrasound Instrument  with a 5–12 MHz linear array transducer. One experienced sonographer collected all images. Images were digitized from the video output of the ultrasound machine using a frame grabber under control of custom software on a personal computer. Image acquisition was gated with an ECG signal so that images were captured at end diastole in each cardiac cycle.Brachial artery diameter was assessed by a single trained operator using an automated, beat-by-beat image processing software package . This analysis technique was previously introduced  and indeBL) measured in millimeters; peak FMD measured in percentage and calculated as [(Dmax-DBL)/DBL] × 100; time-to-peak FMD measured in seconds; diameter change at the end of the post-cuff deflation period (Dend) measured in percentage and calculated from the average diameter of the last 10 seconds of the post-cuff release period relative to DBL.Custom software developed in the MATLAB programming environment  was used to extract measurements from the time series of diameter measurements. The capture time for each image was decoded by optical character recognition of the time of day displayed on each image frame. The sequence of diameter measurements was then interpolated to equally-spaced time samples at 1-second intervals. This interpolated sequence was smoothed by a 9-point median filter followed by a 9-point Gaussian-weighted moving average. The following parameters were measured from the resulting time sequence of brachial artery diameter .Descriptive data are presented as mean ± standard error (SE) unless specified otherwise. Dt test.Variability and repeatability of the test were measured and compared between the two cuff positions. Assessment of variability included: calculation of the coefficient of variation (CV) of the baseline diameters from accepted frames to determine the within-session variability; mean CV from 2 or more baseline measurements for each of the first 19 participants as a day-to-day variability. Both descriptors were calculated for the two cuff locations and then compared by independent Repeatability for FMD analysis for each cuff location was calculated using intraclass correlation coefficient (ICC) in a two-way mixed effects model, and Bland & Altman plots with calculation of the repeatability coefficient. SPSS 12p <0.05.In addition, a linear model was run to assess whether the absolute values of the difference in FMD between test and retest were different using the proximal versus the distal circulatory occlusion. To account for the correlation between distal and proximal comparisons done on same subjects, a generalized estimating equations (GEE) approach was used, assuming an exchangeable correlation structure. S-plus 7.0 (Insightful Corp. 2005) statistical software was used. Results from all tests were considered statistically significant if A total of 62 ultrasound examinations of the brachial artery followed by assessment of endothelium-dependent flow mediated vasodilation were analyzed in this paper, 37 with upper arm cuff location, and 25 with forearm cuff location. Participants' mean age was 28.4 . More male (N = 16) than female (N = 10) participated in the study.p = 0.058). This difference between baseline diameters appears to be related to the slightly different location of the ultrasound transducer on the participants' arm. In fact, similar results were obtained after limiting analysis to those participants that were evaluated with both proximal and distal occlusion (data not shown). Mean number of frames obtained for baseline diameter assessment was similar for both proximal and distal cuff locations (81.7 vs. 81.5 respectively).Overall, mean baseline brachial artery diameter from 37 sessions with upper arm occlusion was smaller (3.8 ± 0.1) than that from 25 sessions acquired with forearm occlusion (4.2 ± 0.2)  was greater than that for the forearm (N = 25) . The gender and the pressure to which the cuff was inflated did not alter these results (data not shown).The time course of the average dilation of the brachial artery from cuff release is shown in figure p = 0.0001). After peak dilation, the artery diameter decrements were slower with proximal occlusion than distal occlusion. At the end of the three-minute image capture period after cuff release, Dend was 11.3 ± 1.2% with proximal cuff occlusion, and 3.3 ± 0.8% with the distal cuff occlusion (p = 0.0001).On average, the time-to-peak FMD was 107.2 ± 6 seconds for upper arm cuff inflation and 67.8 ± 8.9 seconds for forearm cuff inflation  grouped in a cumulative frequency representation shows that utilizing the proximal arm test, ~35% of subjects had not reached their maximal FMD response by the average time (107.2 seconds) Figure . For theFor the 62 studies for which ~80 baseline frames were taken in a single session, the within-session diameter variability was small (CV = 0.85 ± 0.04%); there was no statistically significant difference between the upper arm cuff location (CV= 0.85 ± 0.05%) and the more distal location (CV= 0.84 ± 0.06%). Day-to-day variability of baseline diameter was also small for both the distal cuff placement (CV = 3.4 ± 1.3%) and the proximal one (CV = 5.1 ± 1.7%), with no statistically significant difference between them.The repeatability of the test for peak FMD using Bland-Altman plots is presented in figure p = 0.013) greater when the measurements were done using the proximal method as compared to the distal method.The intraclass correlation coefficient (ICC) between peak-FMD of two repeated measures was 0.6 (95%CI -0.2:0.8) for the upper arm occlusions, and 0.6 (95%CI -0.7:0.9) for the forearm occlusion. Using the GEE approach we found that the magnitude of the differences between the repeated measures were on average 2.4% , or distal (forearm) occlusion has been found to be correlated with coronary endothelial function,13, we eWe have confirmed the well established finding that the maximal FMD is greater and delayed with upper arm occlusion,14-20. WThe time to maximal FMD may vary not only by occlusion site, but also by the method used to detect the diameter from captured frames. For example, adhering to the recommended time interval to maximal FMD at 60 seconds after the release of the occlusive cuff  would hain vivo endothelial function in humans. It has been stated that the primary mechanism responsible for vessel dilation with forearm occlusion is endothelial release of nitric oxide(NO)[end 11.3% vs. 3.3% p <0.0001) for proximal occlusion relative to distal occlusion may reflect other underlying mechanisms. NO is not the only mediator involved in FMD following proximal and distal occlusions; other endothelium-derived factors are probably involved differently with the two occlusion locations. It is essential to conduct validation studies in order to understand the role of different components of endothelial function in one approach or the other of the test.The assessment of FMD of the brachial artery is a functional bioassay for oxide(NO). Doshi eoxide(NO). Additiooxide(NO),25. TherMany studies have assessed the variability of the brachial artery diameter measurements, both at baseline and during the dilatory response. While the baseline diameter was shown to have relatively small variability (CV range 1–13%), the maximal FMD was associated with a much greater range of variability (CV range 1–84%). VariousIn this study we did not quantify the shear stress. It has been observed that the shear stress computed as area under the curve, is the major contributor to the magnitude of FMD. TherefoWe also assessed the repeatability of FMD, an important quality of this outcome as a potential tool in clinical studies where repeated measures are needed, and as a potential indicator for the risk of future coronary disease. Previous studies have shown inconsistent results of brachial artery FMD repeatability. Malik at al reportedFor brachial artery reactivity to be a widespread clinical and research tool, the flow mediated dilation measurement of the brachial artery should be standardized, and centers should adhere to a more consistent analysis protocol. Our data suggest that the choice of the location of cuff inflation may have statistical implications that favor the forearm approach. There are established physiological reasons that support employing one approach over the other. While FMD following distal occlusion is mainly NO dependent, other – not as well elucidated – factors are involved in the dilatory response to upper arm occlusion. In future studies we will study the effects of an environmental insult on FMD of the brachial artery, while measuring different components of endothelial function. In the absence of a well characterized mechanism involved in the flow mediated dilation of the brachial artery with upper arm occlusion, based on our results, we suggest that the forearm occlusion be used. The cuff should be inflated five cm distal to the elbow for five minutes, then, in the post-hyperemic phase of the test the brachial artery should be scanned continuously. Post-cuff deflation examination should be continued for three minutes in healthy adults, whereas, in adults at risk for CAD or with known disease, it is likely necessary to capture the artery's images for longer periods. Last, we suggest also that the images be analyzed with an automated computer-based edge detection algorithm. This will allow for a time series analysis of all brachial artery diameters and identification of the true maximal FMD. More studies performed with standardized method are needed to confirm our findings and allow for potential future use of the test for clinical purposes.The author(s) declare that they have no competing interests.AP – participated in data collection, performed analysis and interpretation of data and drafting of the manuscript. DL – participated in analysis of data and drafting of the manuscript. JS – contributed to conceive the study and its design and participated in interpretation of data. CT – participated in the design of the study and its coordination. FS – participated in data collection and analysis. MA – carried out coordination of the study. MP – participated in the design of the study and data collection. EG – participated in the design of the study and interpretation of data. JK – conceived the study and participated in its design and interpretation of data. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


In Canada – a low endemicity country, vaccines for hepatitis A virus (HAV) are currently recommended to individuals at increased risk for infection or its complications. Applying these recommendations is difficult because the epidemiology of HAV infection is poorly defined, complex, and changing. This systematic review aimed to 1) estimate age-specific prevalence of HAV antibody in Canada and 2) evaluate infection-associated risk factors.MEDLINE (1966–2005) and EMBASE (1980–2005) were searched to identify relevant studies for the systematic review. Archives for the Canada Diseases Weekly Report (1975–1991) and Canada Communicable Disease Report (1992–2005) were searched for relevant public health reports. Data were abstracted for study and participants' characteristics, age-specific prevalence, and risk factors.A total of 36 reports describing 34 unique studies were included.The seroprevalence in Canadian-born children was approximately 1% in ages 8–13, 1–6% in 20–24, 10% in 25–29, 17% in 30–39, and increased subsequently. In age groups below 20 and 20–29, age-specific seroprevalence generally remained constant for studies conducted across geographic areas and over time.Compared to Canadian-born individuals, subjects born outside Canada were approximately 6 times more likely to be seropositive . Travel to high risk areas in individuals aged 20–39 was associated with a significant increase in anti-HAV seropositivity . Compared to heterosexuals, men having sex with men were only at a marginally higher risk . High risk for seropositivity was also observed for Canadian First Nations and Inuit populations.Results from the current systematic review show that in this low endemicity country, disease acquisition occurs in adulthood rather than childhood. The burden of disease is high; approximately 1 in 10 Canadians had been infected by ages 24–29. The increase in prevalence in young adults coincides with disease importation and increasing frequency of risk factors, most likely behavioral-related ones.Gaps in seroprevalence data were identified rendering the application of current immunization recommendations difficult. A nationwide prevalence survey for all Canadians is needed. This is essential to quantify the effectiveness of current recommendations and conduct cost-effectiveness evaluations of alternative immunization programs, if necessary. Hepatitis A virus (HAV) is prominent in many areas of the world -3. In NoApplying the Canadian recommendations is difficult because the epidemiology of HAV infection in Canada is poorly defined, complex, and changing . ReporteEvaluating the impact of the current recommendations is also difficult. Data are needed to distinguish between a cyclical decline and a further decline associated with the recommendations . Such asCase-notification data is of limited use due to under-detection of sub-clinical infection and under-reporting of confirmed cases ,14. A usMEDLINE (Jan. 1966 – Mar. 2005) and EMBASE (Jan.1980 – Mar. 2005) were searched to identify citations of potentially relevant studies for the systematic review (MeSH terms: "hepatitis" exploded AND "Canada" exploded). A study report was included if it contained prevalence data of HAV-antibody  for a Canadian population. Reasons for exclusion were categorized and reported. Citations were screened independently by two reviewers. Full-text study reports from citations deemed relevant by one reviewer were obtained. Archives for the Canada Diseases Weekly Report  and Canada Communicable Disease Report (Jan. 1992 – Mar. 2005) were also searched for potentially relevant public health reports  according to a national survey conducted in 2003 . The serThere was no association between seropositivity and gender based on 9 population comparisons from 5 studies  ,35,41,54Compared to Canadian-born individuals, subjects born outside Canada were approximately 6 times more likely to be seropositive (n = 3008, URR: 5.7 , Table Travel-related data were available in 6 studies Table  23,35,4,49,53,54Two studies evaluated HAV infection among MSM in two different cities ,28. MSM Data on street-involved populations were available in three studies Table  25,29,3,329,30. Seroprevalence in Canadian First Nations and Inuit populations were reported in four studies and summarized in a systematic review . The preAmong Canadian-born day-care educators, there was a borderline significant association between risk of HAV positivity and years of employment Table  23]. A . A 23]. Other potential risk factors were also examined. From a sample of 343 individuals who tested positive for hepatitis C, 30% of those aged 20–29 were seropositive for HAV . One proThe seroprevalence data consolidated in this systematic review had many limitations. Except for one national survey in ages 8–13 , other sResults from the current systematic review show that disease acquisition occurs in adulthood rather than childhood . In CanaA low level of HAV immunity in Canada is evident from this systematic review. Over 90% of Canadian-born individuals aged 20–29, and over 80% of those aged 30–39 remained unprotected. Canadians born outside the country generally have a higher prevalence of HAV antibody, yet including these individuals did not significantly improve the percentage of protected Canadians. This low level of immunity and persistent risk of exposure to HAV suggest that outbreaks are possible in the future ,47. For Clarifications are required to better understand the epidemiology of HAV in Canada, especially the inter-relation between timely case-notification data, seroprevalence data, and risk factor data. In a study examining national case-notification data from 1990–1999, estimated incidence of reported cases decreased while the average age of exposure and subsequent infection increased . Given tResults of the current systematic review are consistent with low HAV endemicity patterns in developed countries -62. A suIn order to apply the current immunization recommendations, substantial information pertaining to groups at increased risk of HA infection or its complications is required. Results from this systematic review suggest that the risk of HA infection in these target groups was not well documented. With the exception of a few population-based surveys ,49, mostResults from the current systematic review show that in this low endemicity country, disease acquisition occurs in adulthood rather than childhood. The burden of disease is high; approximately 1 in 10 Canadians had been infected by ages 24–29. The increase in prevalence in young adults coincides with disease importation and increasing frequency of risk factors, most likely behavioral-related ones.Gaps in seroprevalence data were also identified in this systematic review, rendering the application of current recommendations difficult. A nationwide prevalence survey for all Canadians is needed. This is essential to quantify the effectiveness of current recommendations  and condG6 students grade 6 studentsFunding for this systematic review was provided by GlaxoSmithKline Canada. BP and ACT are employed by GlaxoSmithKline. BD, GDS, VG, JO, and DS have received research funding from GlaxoSmithKline.BP contributed to the development of the research question and methodology, project management, systematic review, data management and data analysis, derivation of charts and tables, and interpretation of the results. BD, GDS, DS contributed to the development of the research question and methodology, acquisition of unpublished data, and interpretation of the results. JO and VG contributed to the development of the research methodology, acquisition of unpublished data, derivation of charts and tables and interpretation of the results. ACT contributed to the development of the research methodology, project management, systematic review, derivation charts and tables and interpretation of the results. All of us contributed to the manuscript writing and approved the final version of the manuscript.The pre-publication history for this paper can be accessed here:


Epidemiological studies may be subject to selective reporting, but empirical evidence thereof is limited. We empirically evaluated the extent of selection of significant results and large effect sizes in a large sample of recent articles.p = 0.25). Paradoxically, the smallest presented relative risks were based on the contrasts of extreme quintiles; on average, the relative risk magnitude was 1.41-, 1.42-, and 1.36-fold larger in contrasts of extreme quartiles, extreme tertiles, and above-versus-below median values, respectively (p < 0.001).We evaluated 389 articles of epidemiological studies that reported, in their respective abstracts, at least one relative risk for a continuous risk factor in contrasts based on median, tertile, quartile, or quintile categorizations. We examined the proportion and correlates of reporting statistically significant and nonsignificant results in the abstract and whether the magnitude of the relative risks presented (coined to be consistently ≥1.00) differs depending on the type of contrast used for the risk factor. In 342 articles (87.9%), ≥1 statistically significant relative risk was reported in the abstract, while only 169 articles (43.4%) reported ≥1 statistically nonsignificant relative risk in the abstract. Reporting of statistically significant results was more common with structured abstracts, and was less common in US-based studies and in cancer outcomes. Among 50 randomly selected articles in which the full text was examined, a median of nine (interquartile range 5–16) statistically significant and six (interquartile range 3–16) statistically nonsignificant relative risks were presented . These studies have shown that very many of the findings of trials are never published, and that statistically significant findings are more likely to be included in published papers than nonsignificant findings. However, much medical research is carried out that does not use randomized trial methods, either because that method is not useful to answer the question at hand or is unethical. Epidemiological research is often concerned with looking at links between risk factors and the development of disease, and this type of research would generally use observation rather than experiment to uncover connections. The researchers here were concerned that selective reporting bias might be just as much of a problem in epidemiological research as in randomized trials research, and wanted to study this specifically.In this investigation, searches were carried out of PubMed, a database of biomedical research studies, to extract epidemiological studies that were published between January 2004 and October 2005. The researchers wanted to specifically look at studies reporting the effect of continuous risk factors and their effect on health or disease outcomes . Three hundred and eighty-nine original research studies were found, and the researchers pulled out from the abstracts and full text of these papers the relative risks that were reported along with the results of statistical tests for them.  The researchers found that nearly 90% of these studies had one or more statistically significant risks reported in the abstract, but only 43% reported one or more risks that were not statistically significant. When looking at all of the findings reported anywhere in the full text for 50 of these studies, the researchers saw that papers overall reported more statistically significant risks than nonsignificant risks. Finally, it seemed that in the set of papers studied here, the way in which statistical analyses were done produced a bias towards more extreme findings: for datasets showing small relative risks, papers were more likely to report a comparison between extreme subsets of the data so as to report larger relative risks.These findings suggest that there is a tendency among epidemiology researchers to highlight statistically significant findings and to avoid highlighting nonsignificant findings in their research papers. This behavior may be a problem, because many of these significant findings could in future turn out to be “false positives.” At present, registers exist for researchers to describe ongoing clinical trials, and to set out the outcomes that they plan to analyze for those trials. These registers will go some way towards addressing some of the problems described here, but only for clinical trials research. Registers do not yet exist for epidemiological studies, and therefore it is important that researchers and readers are aware of and cautious about the problem of selective reporting in epidemiological research.http://dx.doi.org/10.1371/journal.pmed.0040079.Please access these Web sites via the online version of this summary at publication bias (note: Wikipedia is an internet encyclopedia that anyone can edit)Wikipedia entry on International Committee of Medical Journal Editors gives guidelines for submitting manuscripts to its member journals, and includes comments about registration of ongoing studies and the obligation to publish negative studiesThe ClinicalTrials.gov and the ISRCTN register are two registries of ongoing clinical trials Researchers sometimes selectively present their findings, focusing on the more impressive aspects of their work. Focusing on impressive aspects means that researchers may try to show statistically significant results and/or larger effect sizes. Testing for statistical significance is not necessarily a bad thing, even though the process has been criticized . In theoWhile these biases have been studied predominantly in the randomized trials literature, selective reporting may be more prominent in epidemiological research . MoreoveMoreover, one may examine how epidemiological estimates of risk are presented in the literature. Is there a preference to show larger magnitudes of effect? For risk factors that take continuous values, the presented magnitude of the relative risk may depend on the selected contrast. The risk may be presented per unit change, per standard deviation, or according to a contrast of specific percentile groups of the values of the postulated risk factor. The latter option is very popular. Typical contrasts involve splitting the data into quintiles, quartiles, tertiles, and above-versus-below the median value of the risk factor. Unless a risk relationship is J- or U-shaped, when the compared groups are further apart, the estimated relative risk would deviate further from the “null” value of 1.00. For example, if consumption of a nutrient is associated with the risk of prostate cancer, then the risk ratio may be 1.10 when values of above-versus-below the median are compared, but in the same data, the risk ratio may be 1.50 if extreme quintiles are compared. When relative risks are inherently more modest, do investigators select to compare groups that are further apart?Here we examined empirically a sample of published articles on epidemiological studies that presented relative risk estimates for continuous risk factors with percentile-based contrasts. The analysis had two objectives. First, we aimed to estimate how many articles highlight “positive” or “negative” results and to identify correlates thereof. Second, it was our intention to identify whether the magnitude of the highlighted risk was related to the selected percentile contrast for the risk factor.We assembled a systematic sample of recent epidemiological articles that presented, in their respective abstracts, at least one relative risk for at least one continuous risk factor in contrasts based on median, tertile, quartile, or quintile categorizations. The following seven contrasts were considered eligible: above-versus-below median values; extreme tertiles; extreme tertile versus remaining subjects; extreme quartiles; extreme quartile versus remaining subjects; extreme quintiles; and extreme quintile versus remaining subjects. A pilot literature screen suggested that other percentile-based categorizations are uncommon. Contrasts between two extreme groups were considered eligible, regardless of whether estimates were also presented for contrasts between one extreme and one intermediate group. For example, the extreme quartiles category considers all studies in which a relative risk for the contrast of extreme quartiles is presented, regardless of whether data are presented for the contrast of the third versus first quartile and the second versus first quartile. Whenever the contrast was not clear from the abstract, we searched the full article. We did not consider slope estimates, i.e., a change in odds ratio for each unit change in exposure, since this does not require any choice for categorization of exposure levels.http://www.ncbi.nlm.nih.gov/entrez/query.fcgi?DB=pubmed), combining the term cohort* with quintile*, quartile*, tertile*, or (median AND [above OR below]) as words in the abstract. Only English-language articles were considered. We included all eligible retrieved articles published in a period starting 1 January 2004 and indexed until the last search update (28 October 2005).We accepted all multiplicative-effect metrics as measures of relative risk, including hazard ratios, incidence rate ratios, risk ratios, and odds ratios. We did not consider studies where a continuous outcome, rather than the postulated risk factor, was categorized according to some percentile grouping. We considered epidemiological studies regardless of their design, but excluded meta-analyses, as we focused on primary studies. We used a search strategy that would favor the selection of cohort studies. Cohort studies are traditionally the most definitive type of epidemiological investigation. However, they are less common in the literature than case-control and cross-sectional designs. To enrich our sample with cohort designs, we searched PubMed  or 95% confidence intervals (CIs).For each article that reported, in the abstract, at least one numerical relative risk estimate for any of the seven eligible percentile-based contrasts, we recorded the first author, journal of publication, impact factor of the journal , first We also extracted detailed data for one eligible relative risk per study. Epidemiological studies often test numerous hypotheses and estimate numerous relative risks. Investigators may mention in the abstract what they consider to be the key findings of their work. To avoid subjective selection on our part, we always chose the first eligible relative risk presented numerically in the abstract. The choice of the first presented relative risk has been adopted also in previous empirical research on epidemiological studies . For thiRisk factors were categorized as dietary intake and dietary patterns; toxic exposures and markers thereof; biological markers in any biological fluid or tissue; psychological, behavioral, or social factors; physical activity or energy expenditure; body characteristics and composition; and other. Outcomes were classified as mortality versus non-mortality; the latter were further classified into malignancies, vascular , and other.We did not record the sample size of each study, because the pertinent sample for each presented relative risk might differ from the overall sample. Instead, we calculated the standard error of the natural logarithm of the relative risk. The standard error is given by the absolute value of the difference between upper and lower 95% CI divided by 3.92. The inverse of the standard error is a more appropriate measure of precision than plain sample size.Abstracts may be biased towards reporting the most significant results, but such selection may not affect the information available in the full text. Moreover, another issue concerns whether the predominance of significant results in the abstract might be less prominent if all presented relative risks were examined rather than the first one alone. Therefore, for sensitivity analyses, we also evaluated 50 randomly selected articles in more depth, where we recorded all the respective relative risks presented in both the abstract and in the full text.All data were extracted in duplicate by two independent investigators. Discrepancies were resolved by consensus.p < 0.25 in the univariate analysis; we then used backward elimination using likelihood ratio criteria with a p = 0.05 threshold. All analyses that we performed for all variables considered are shown in tabulated form for the odds ratios derived from the regression coefficients and their 95% CIs. No varying forms models or complex interaction terms were evaluated. Similarly, we examined whether there was any relationship between these specific variables and the reporting of at least one statistically nonsignificant relative risk.We evaluated the proportion of studies that had at least one statistically significant relative risk estimate presented numerically in the abstract. We examined whether this was related to any of the study characteristics mentioned above. We first performed univariate logistic regressions. In multivariate models, a variable was considered initially only if it had In sensitivity analyses using full texts (50 articles), we used Wilcoxon rank sum tests to examine whether the number of significant relative risks presented in the abstract and in the full text of results exceeded the nonsignificant ones. We tested (using McNemar and Wilcoxon signed rank tests) whether the proportion of significant relative risks and the magnitude of the relative risks were different, and when the first eligible relative risk presented in the text was selected, rather than the time of selecting the first eligible relative risk presented in the abstract.Analyses of the magnitude of the presented relative risks used only the first eligible relative risk per study presented in the abstract. We coined all relative risk estimates in such a way that they would be ≥1.00; i.e., relative risks <1.00 were inverted so as to focus consistently on the extent of deviation from the “null.” Values were log-transformed. Analysis of variance (ANOVA) compared first the four contrasts of two extreme groups . Then, we considered both types of contrasts  and type of percentile; the above-versus-below median contrast was not relevant in this analysis.Finally, we examined with ANOVA whether the choice of contrast was related to the degree to which a study was well powered, i.e., whether studies chose contrasts comparing more extreme groups, and when they had more sufficient power (lower standard errors of the relative risk estimates).http://www.spss.com). p-Values are two-tailed.Analyses were conducted in SPSS 13.0  [n = 1], JAMA: The Journal of the American Medical Association [n = 7], and New England Journal of Medicine [n = 3], although none appeared in Lancet or BMJ). Almost a quarter of the 389 studies appeared in journals that have an impact factor of above seven. Half of the 389 studies originated from the United States. Twenty-four cohorts contributed more than one article  to the eligible 389 articles. In particular, 54 articles (13.9%) originated from the large cohorts established at Harvard University , followed by ARIC  and EPIC . Typical cohorts were overrepresented, as intended. Structured abstracts were very common. The comparison of extreme quartiles was the most frequent type of contrast among the first presented relative risks. Contrasts of extreme tertiles and extreme quintiles were also common. Contrasts of one extreme group against the remaining subjects were far less frequent.The electronic search yielded 850 articles; 461 were rejected upon screening  and 389 The median was 1.73 (interquartile range [IQR] 1.39–2.48) when all relative risks were coined to be ≥1.00, and only 22 of these relative risks (5.7%) exceeded 5.0. Four-fifths of these relative risks were statistically significant . Among rAlmost nine out of ten articles  reported ≥1 statistically significant relative risk in the abstract. The proportion of abstracts with ≥1 statistically nonsignificant relative risk was only 169/389 (43.4%). Among 30 excluded articles where otherwise eligible relative risks appeared, but without exact numerical information, proportions were 93.3% and 56.7%, respectively.Reporting at least one statistically significant risk in the abstract was positively related to the presence of a structured abstract, and it was less likely in studies originating from the United States and in studies of cancer outcomes in both univariate and multivariate analyses . Conversp < 0.001).In the 50 randomly selected articles evaluated in depth, the abstracts had a median of two statistically significant relative risks (IQR 1–3) versus zero nonsignificant risks (IQR 0–1). A preponderance of statistically significant relative risks was seen in 33 articles (66%), a preponderance of nonsignificant risks was noted in five articles (10%), and equal numbers were seen in 12 articles (24%) . None of the 50 articles reported only a single relative risk. Only four out of the 50 articles claimed to be the first study assessing a specific association, while a further five articles claimed to be the first to address a previously tested association in a new setting (based on population characteristics or type of study design).In the full text, the median number of articles reporting any statistically significant relative risk was nine (IQR 5–16) versus six articles reporting nonsignificant risks (IQR 3–16). In 28 articles (56%), more statistically significant relative risks were reported than statistically nonsignificant relative risks. In 20 articles (40%), there were more nonsignificant relative risks than significant ones, and in two articles (4%) there was an equal number of significant and nonsignificant relative risks .The first reported relative risk with eligible contrast in the full text was formally statistically significant in 35 articles (70%). The median relative risk was 1.54 (IQR 1.30–2.43), when relative risks were consistently coined to be ≥1. These values are similar to the respective values for the first reported relative risk with eligible contrast in the abstract  was not seen in any of these 50 articles.p < 0.001).p = 0.044) for comparisons of the two extreme groups versus comparisons of one extreme group against the remaining subjects after adjusting for type of percentile involved in the contrast . Only six studies referred explicitly to a J- or U-shaped or nonlinear effect. Exclusion of these studies did not change our results (unpublished data).Moreover, the presented effects were smaller, on average, when extreme groups were compared than when one extreme group was compared against the remaining subjects . The relp < 0.001). Thus, precision was maximal when the compared extreme groups pertained to the smallest possible portion  of the study population.Compared with contrasts of above-versus-below the median, the standard error of the relative risks was similar, on average, in studies with contrasts of extreme tertiles (on average 1% larger) or extreme quartiles . The standard error was 49% lower on average when extreme quintiles were contrasted  that arises out of a single hypothesis and a single analysis is very different than one that arose out of a massive screening of potential risk factors where it is not shown that many other risk factors have also been screened.We should also acknowledge that researchers and editors may try to select and present what they deem to be the most interesting and important work. The window on the world offered by the published scientific literature is not comprehensive, but instead is a particular view reflecting a host of complicated desires, abilities, and interests of the scientific community. Whether statistical significance should be one of the criteria used to select work for presentation has been a point of endless debate. However, at a minimum, if data are selected based on significance thresholds, it is important to know the underlying multiplicity of the conducted analyses. A significant risk ; preliminary evidence suggests that the pursuit of statistical significance exists across all epidemiological studies . Second,Selection in primary studies could also affect the findings and inferences of secondary analyses, leading to spurious conclusions being drawn. A meta-analysis may ameliorate this defect by using the raw  data without taking into consideration the selected contrasts that had been presented in each paper. However, this would require full access to all data, and this is currently the exception. Selection of outcomes and contrasts in the primary studies may lead to similar selective choices in meta-analyses that have to depend on published data. This would further perpetuate these biases. Meta-analyses may try to detect and address these selective reporting problems using a variety of diagnostic tools such as asymmetry tests. However, having an unbiased body of evidence is certainly preferable to trying to detect and eliminate bias after the fact.Our empirical findings may lead to some recommendations on how to improve the situation. Epidemiological research is very important , but repTable S1(162 KB PDF)Click here for additional data file.


To identify risk factors for sudden infant death syndrome relevant in Lithuania.A nationwide case-control study surveying parents of 35 infants who died from sudden infant death syndrome during the period of 1997–2000 and parents of 145 control infants matched with SIDS infants for date of birth and for region of birth was carried out.vs. cold period (40%). Prone and side sleeping positions both carried no increased risk of sudden infant death syndrome compared with supine because of a rare prone sleeping (4.1% of controls vs. 0% of dead infants) and more prevalent side than supine sleeping (84.8% of controls vs. 94.3% of dead infants) in the controls as well as the cases. Bed sharing for the whole night as a risk factor for sudden infant death syndrome has not been confirmed, either, as bed sharing was common only for the controls (13.8% of controls vs. 0% of dead infants). Routine sleeping environment factors such as heavy wrapping (≥4 togs) of an infant , sleeping in a bassinet  and maternal factors such as maternal education ≤12 years , unplanned pregnancy  and ≥2 previous live births  were significantly associated with sudden infant death syndrome on multivariate analysis.Deaths incidence was greater in the warm period (60%) The results of this first population-based case-control study have shed some light on the epidemiology of the syndrome in Lithuania. Although the mortality of sudden infant death syndrome in Lithuania is not high, it might be lowered moreover by public informing about sudden infant death syndrome and related risk factors. Special attention must be paid to mothers with low education on potentially modifiable risk factors such as routine heavy wrapping of an infant during sleep, routine sleeping in a bassinet and unplanned pregnancy. Despite the fact that sudden infant death syndrome (SIDS) is the leading cause of postneonatal infant mortality in most developed countries, SIDS incidence varies greatly in different countries and between regions within countries worldwide. Before 1990–1991, SIDS incidence varied from 1 to 6 cases per 1000 live births . UnfortuAs considerable differences of SIDS risk factors importance may exist in different countries, the aim of our study was to identify factors associated with and predicting increased risk of SIDS relevant in Lithuania.The survey was performed as a retrospective case-control study during the period of 2002–2003. The main case-control study instruments were questionnaires for cases and controls. The questionnaire for cases consisted of 89 standardized questions concerning infant death circumstances, demographic factors, routine practices in sleeping environment, infant and maternal medical history, parental socioeconomic status and lifestyle. The questionnaire for controls consisted of 81 standardized questions concerning demographic factors, routine practices in sleeping environment, infant and maternal medical history, parental socioeconomic status and lifestyle. No reference sleep was assigned for the control group as the time frame was too long. Questions were of a "yes" or "no" or multiple choice nature or needed data in figures, such as birth date, birth weight, death time, and others.Questionnaires for cases were completed by the research interviewer during a home visit. Dialogue about aims of visit took the priority. Afterward standardized questions for SIDS cases were asked. SIDS parents were questioned only on a receipt of underwritten consent of participation in the study. The mean time between SIDS death and completing the questionnaire was 3.9 ± 0.2 years. Questionnaires for controls together with a sheet of information and consent of participation in the study were distributed by mail to control parents.Data of SIDS infants were obtained from computerized database of Lithuanian Department of Statistics. Initially all of 45 infants whose death was attributed to sudden infant death syndrome (ICD-10 – R 95 and R 96) during the period of 1997–2000 were included in the study. Data of control infants were obtained from Lithuanian Medical Birth Registry. Initially 225 controls matched with SIDS infants for date of birth within one month and for region of birth were picked randomly. Selection results were rectified according to Lithuanian Infant Mortality Registry data and 3 infants who died during the period of 1997–2000 were excluded from the control sample.Statistical Package for Social Science) software.Data of the study were processed using STATISTIKA/w 5 and SPSS/w 10 (Student t and Fisher's (exact) criteria were used for comparison of means. As using these criteria is possible only when sample variables are distributed normally, Kolmogorov-Smirnov goodness-of-fit test was performed in order to assess the normality of distribution.Chi-square and logistic regression analyses were used to examine differences in the prevalence and prediction of various sleeping environment, maternal, infant, parental socioeconomic and lifestyle risk factors among SIDS cases and controls. Fisher's exact test was utilized for small cell frequencies. Variables significant on univariate analysis were included in multivariate analysis using a conditional logistic regression. Later on, multivariate models were constructed with the backward stepwise procedure for variables significant at the 5% level. The estimations of both univariate and multivariate analysis resulted in odds ratio (OR) with their 95% confidence intervals (CI).The hypotheses were considered statistically significant at the level of p < 0.05, very significant at the level of p < 0.01 and especially significant at the level of p < 0.001. The hypotheses were considered statistically not significant at the level of p > 0.05.10 SIDS cases (22.2%) of the total 45 were excluded from the analysis: 5 because the families could not be traced, having moved from the notified inhabitation place; 2 because the families refused an interview and 3 because they fell short of accepted SIDS definition criteria after review of the case history on the grounds of an interview. We defined SIDS according to Willinger as the sudden death of an infant aged from 7 to 365 days, which remained unexplained after performance of a complete postmortem investigation, including an autopsy, examination of the death scene, and review of the case history [Finally, data of 35 SIDS cases and 145 controls (1:4) have entered the final statistical analysis.2 test of the null hypothesis of a uniform frequency distribution over 12 months was significant . The proportion of SIDS deaths was greater in the first half-year of age (82.8%) than in the second half-year (17.2%) with a peak incidence from 2 to 4 month of age (59.9%)  age at death in SIDS infants was 114.1 ± 9.6 days and varied from 23 to 235 days. A χvs. cold period (40%).SIDS deaths according to season were classified into two periods – cold period (six lowest temperature months – October to March) and warm period (six highest temperature months – April to September) Figure . SIDS deMost of the SIDS deaths occurred during what the parents classified as night sleep. 74.3% of infants (26/35) were discovered dead between 1.50 and 9.00 hours and 25.7% (9/35) – between 9.30 and 14.30 hours. The interval between the time that infants were last seen or heard alive and the time found dead was 186.2 ± 24.6 min.A univariate analysis was carried out to estimate the strength of the relationship between SIDS and factors concerning routine practices in sleeping environment, infant and maternal medical history, parental socioeconomic status and lifestyle. Within view of SIDS mothers reluctance to report alcohol consumption, especially frequency of alcohol consumption, this variable had not been involved into univariate analysis. Frequencies, odds ratios and p values of all the univariate findings are presented in Table Over 30 factors were analyzed, and more than half of them were significant. Factors significant on univariate analysis were routine sleeping in a bassinet, routine use of substandard infant mattresses, such as pallet, twisted plaid or pillow for sleep, routine use of waterproof cloth over the mattress for sleep, routine wearing of ≥4 togs during sleep, routine wearing of cap during sleep, routine bed sharing with parent(s) for the whole night, birth weight <2500.0 g, gestation ≤36 weeks, bottle feeding off the birth, unplanned pregnancy, late  or no prenatal care, ≥2 previous live births, maternal age ≤20 or ≥35 years, paternal age ≤20 or ≥35 years, maternal education ≤12 years, single mother , households without waged income, worse than fair housing conditions, worse than fair self-perceived economic family status, overcrowding (>1 person per room), maternal smoking during pregnancy, paternal smoking during pregnancy, and exposure to smoking during pregnancy when just one or both of the parents smoked.Examples of factors that were not significant on univariate analysis included routine sleeping positions, routine sleeping in a full-sized bed , routinely used bedding type, routine use of pillow, routine sleeping in room alone, male sex, any neonatal problem, any congenital problem, twin birth, no dummy when sleeping, previous stillbirth, previous interruption of pregnancy, previous infant death, paternal education ≤12 years, and households renting or living with parents.A multivariate analysis was carried out to determine which factors were independently significant when controlled for other factors found to be important in the study. In the course of correlation analysis some correlates have been excluded from the multivariate analysis. So, all variables with an expected cell <5 have been excluded from the multivariate analysis. Odds ratios and p values for factors that remained significant or not significant after controlling for other significant on univariate analysis factors are presented in Tables Table th step  and back (17%) positions are commonly recommended sleep positions at discharge from maternity unit in Lithuania. Epidemiological studies from England and New Zealand have shown that side sleeping has a slightly higher risk of SIDS than the supine position, though not as great as prone sleeping 3% and ba. The higOther results from the sleep surface analysis were unexpected for us, too. The risk associated with routine sleeping in a bassinet, a substandard infant mattress for sleep and a waterproof cloth over the mattress during sleep were significant among the factors associated with the sleeping environment. Routine sleeping in a bassinet and waterproof cloth over the mattress have entered in the multivariate model with the best prognostic power. Sleeping in bassinets, which are designed mainly for carriage, unlike cots, which are designed for sleep and meet safety standards for infants, may, at least in theory, carry a risk of accidental entrapment and suffocation. As we have found only a low correlation between placing an infant to sleep in a bassinet and households having no waged income, this practice is not a marker for a low economic status, but a stereotype of a routine infant sleeping environment in Lithuania. A waterproof cloth over the mattress during sleep as well as other loose bedding may carry a risk of accidental entrapment and suffocation, too. Epidemiological studies worldwide identified soft surfaces, such as pillows, quilts, comforters, sheepskins and porous mattresses as a significant risk factor, particularly when placed under a sleeping infant ,10. HoweSome studies investigating SIDS have reported the risk associated with the excessive amount of clothing or bedclothes and high-temperature environments during sleep, particularly for infants lying prone . In our Some scientific studies have demonstrated that bed sharing can alter sleep patterns of mother and baby ,13. ThesThe contribution of artificial feeding to SIDS is not clear and may vary in different communities. On the basis of our results bottle feeding from the birth is not a risk factor for SIDS when controlled for other significant factors on multivariate analysis. However, there are other good reasons to continue promoting breastfeeding.The higher incidence of cases with low birth weight and preterm birth confirms findings of other studies .Unplanned pregnancy and 2 or more previous live births are significant potentially modifiable risk factors for SIDS found in our study. Although these factors do not point to a specific etiology, they suggest that interventions targeted towards family planning methods may theoretically reduce SIDS incidence in Lithuania.The data from univariate analysis related with socioeconomic household status, such as parental age at the birth of an infant, parental education, waged income, housing conditions, number of persons living per room and self-perceived parental economic situation have demonstrated it to be significantly lower in SIDS group than in control. Further, the multivariate analysis showed maternal education ≤12 years to be one of the very significant independent SIDS risk factors in Lithuania. Mothers with lower education might care differently for their babies than mothers having higher education, and they may possibly have many worries, other than those about their child. A link between low socioeconomic status and SIDS has been noted in the literature .Maternal smoking during pregnancy is a major, potentially modifiable risk factor found in many other studies ,17,18. EThe variables found to be significant in case-control study depend on what is included in a multivariate model. Further more extensive and detailed epidemiological research of SIDS is needed in Lithuania in order to understand the complex relationships between these variables and other factors that affect infant health and serve to heighten the risk of SIDS.The results of this first population-based case-control study of the SIDS in Lithuania have shed some light on the epidemiology of the syndrome in Lithuania. The circumstance that some of these factors may be modifiable has important implications in terms of social policy and health education. Although the mortality of SIDS in Lithuania is not high, it might be lowered moreover by public informing about SIDS and related risk factors. Special attention must be paid to mothers with low education on potentially modifiable risk factors such as routine heavy wrapping of an infant during sleep, routine sleeping in a bassinet and unplanned pregnancy.The author(s) declare that they have no competing interests.The study was conceived and designed by Vilija Bubnaitienė and Ramunė Kalėdienė with assistance from Rimantas Kėvalas. The study was conducted under the supervision of Ramunė Kalėdienė. Statistical analysis was conducted by Vilija Bubnaitienė with assistance from Ramunė Kalėdienė and Irena Nedzelskienė. Interpretation of results was conducted by Vilija Bubnaitienė with assistance from Ramunė Kalėdienė. The manuscript was prepared by Vilija Bubnaitienė and edited by Ramunė Kalėdienė and Rimantas Kėvalas. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here: