Estimation of between-study heterogeneity is problematic in small meta-analyses. Bayesian meta-analysis is beneficial because it allows incorporation of external evidence on heterogeneity. To facilitate this, we provide empirical evidence on the likely heterogeneity between studies in meta-analyses relating to specific research settings.Our analyses included 6,492 continuous-outcome meta-analyses within the Cochrane Database of Systematic Reviews. We investigated the influence of meta-analysis settings on heterogeneity by modeling study data from all meta-analyses on the standardized mean difference scale. Meta-analysis setting was described according to outcome type, intervention comparison type, and medical area. Predictive distributions for between-study variance expected in future meta-analyses were obtained, which can be used directly as informative priors.Among outcome types, heterogeneity was found to be lowest in meta-analyses of obstetric outcomes. Among intervention comparison types, heterogeneity was lowest in meta-analyses comparing two pharmacologic interventions. Predictive distributions are reported for different settings. In two example meta-analyses, incorporating external evidence led to a more precise heterogeneity estimate.Heterogeneity was influenced by meta-analysis characteristics. Informative priors for between-study variance were derived for each specific setting. Our analyses thus assist the incorporation of realistic prior information into meta-analyses including few studies. The existing literature on heterogeneity in meta-analyses of continuous outcomes is sparse, and so little is known as to what forms a realistic prior distribution for the between-study variance. This article proposes a new set of informative prior distributions for use in specific research areas.•We have demonstrated how an informative prior for heterogeneity can be used in a future meta-analysis. In each of two illustrative examples, incorporation of external information led to more precise estimates for the between-study variance.•In view of the strong associations between meta-analysis characteristics and the extent of heterogeneity observed in our data set, the use of an empirically derived informative prior for heterogeneity in future meta-analyses would be perfectly reasonable.1Policy decision makers are becoming increasingly reliant on the findings from systematic reviews In many meta-analyses, there are few studies available to include, perhaps because the disease is rare or the treatment under assessment is new. Of 22,453 meta-analyses from the Cochrane Database of Systematic Reviews (CDSR), containing at least two studies, just under 75% contained five or fewer studies τ2 for use in future binary outcome meta-analyses on the log odds ratio scale.To facilitate Bayesian meta-analysis with an informative prior for the between-study variance, we provide empirical evidence on the likely extent of heterogeneity in meta-analyses of particular settings, defined by outcome type, types of interventions evaluated, and medical area. Study data from the binary outcome meta-analyses in the CDSR have already been analyzed by Turner et al. Here, we analyze data from a large collection of published continuous-outcome meta-analyses and investigate the influence of meta-analysis characteristics on between-study heterogeneity. We provide predictive distributions for the extent of heterogeneity expected in future continuous-outcome meta-analyses in particular settings. These distributions can be used in new meta-analyses as “off-the-shelf” informative prior distributions for the between-study variance 22.1CDSR is a rich resource of systematic reviews in areas of health care. These reviews have been prepared by the Cochrane Collaboration, with the objective to make the most up-to-date and reliable evidence conveniently available to health care consumers, professionals, and providers Cochrane reviews typically include multiple meta-analyses, which correspond to the comparisons of different pairs of interventions or the assessment of different outcomes within the same research area. For example, a review examining antibiotics could report separate meta-analyses comparing each of several antibiotics against a placebo, with respect to both infection severity and adverse effects. Meta-analyses were included in our analyses if they consisted of data from at least two studies. In some reviews, results from studies eligible for a meta-analysis were available, but no pooled results were published in the Cochrane review. Such data were regarded in the same way as meta-analyses to maximize the amount of information available. The review authors may have decided not to perform a meta-analysis based on the degree of heterogeneity between studies Reviews sometimes present results for several subgroup analyses within meta-analyses. Because we are interested in the overall between-study heterogeneity in a meta-analysis, study results were combined across subgroups. In some reviews, the subgroups presented within a meta-analysis were not mutually exclusive; therefore, we checked for study duplications and used data for only the first occurrence of each study in each meta-analysis All meta-analyses in the original CDSR database have been classified according to the type of outcome, types of interventions involved in the pairwise comparison, and medical specialty, as described in an earlier article For each study measured as a continuous outcome, we have study data consisting of means and standard deviations, together with the number of participants in each intervention group. All meta-analyses have been categorized according to whether the meta-analysis was originally published on the mean difference (MD) or standardized mean difference (SMD) scale.2.2www.jclinepi.com (Section A.1.1).We used hierarchical models to analyze study data from each meta-analysis in the data set, while investigating the influence of meta-analysis characteristics on the extent of between-study heterogeneity. Within each meta-analysis, a random-effects model with normal within-study likelihoods was fitted to continuous outcome data from each study, on the SMD scale. A definition of the SMD is provided in the τ2, without accounting for meta-analysis characteristics as covariates. We contemplated several distributions for τ2 and took forward three distributions into later analyses, adjusting for covariates, based on assessment of goodness of fit. Turner et al. τ2 in binary outcome meta-analyses. We contemplated a log-normal distribution for τ2 in continuous-outcome meta-analyses. Other candidate distributions included the heavier tailed log-t distribution with five degrees of freedom, and also an inverse-gamma distribution, as a conjugate prior for the variance of a normal distribution. Model selection based on the deviance information criterion (DIC) t model for τ2.Many meta-analyses in the data set have been published on the MD scale. Nonetheless, we analyzed all study data using the SMD scale, and we compared the distribution of heterogeneity among observed SMDs for meta-analyses originally analyzed on the SMD scale and meta-analyses originally analyzed on the MD scale. Our analyses initially investigated the distributional form of the between-study heterogeneity variance t distribution with five degrees of freedom for residual variation. As covariates in our regression models, we included indicators for outcome type, type of intervention comparison, and medical area. Within pairwise comparisons, heterogeneity was assumed to vary across meta-analyses, with separate variances for the different outcome types. Heterogeneity was also assumed to vary across pairwise comparisons, with separate variances for each type of intervention comparison. The mathematical form of the model is given in the www.jclinepi.com (Section A.1.2).Across meta-analyses, a hierarchical regression model was fitted to log-transformed values of underlying between-study heterogeneity, assuming a All models were fitted using Markov chain Monte Carlo (MCMC) within the WinBUGS www.jclinepi.com (Section A.1.2). A log-t distribution was fitted to each predictive distribution, using posterior quantities for log  were originally performed on the MD scale, and 21%  were originally performed on the SMD scale. Seven hundred twenty-eight studies (2.4%) have missing standard deviations and are therefore removed from our statistical analysis. Twenty-eight meta-analyses (0.4%) were excluded from our analyses in which the outcome type did not fit into any of our predefined categories and was classified as “other.” Frequencies of outcome types, types of intervention comparison, and medical areas among the remaining 6,492 meta-analyses in our data set are given in τ2 on this scale was negative and hence set to zero. Nonzero estimates for τ2 have a median of 0.10 and 95% range of 0.002–2.30. Among the meta-analyses analyzed originally on the raw MD scale but reanalyzed on the SMD scale, 43% of method of moment–based estimates for τ2 were negative and hence set to zero. Nonzero estimates for τ2 have a comparable median and 2.5% quantile to the meta-analyses analyzed originally on the SMD scale, a median of 0.11 and 95% range of 0.002–4.38. Histograms representing the empirical distributions of nonzero estimates for τ2 on the log scale are provided in the www.jclinepi.com (Section A.2). The distributions based on analyses of MDs and SMDs are broadly similar, and in the remainder of the article, we use the complete data set, analyzed throughout on the SMD scale.In approximately 40% of meta-analyses analyzed originally on the SMD scale, the method-of-moments estimate for 3.2t distribution with five degrees of freedom led to a DIC value of 19,562, compared with 29,565 for the inverse-gamma model and 19,582 for the log-normal model for τ2. Thus, the log-t regression model for τ2 appears to be the better choice. The inverse-gamma model seems a poor fit. In this section, we focus on results from fitting the log-t model to investigate the influence of meta-analysis characteristics on the extent of heterogeneity in a meta-analysis. To compare levels of between-study heterogeneity across different meta-analysis types, we report ratios of heterogeneity variances τ2, together with their respective 95% credible intervals (CIs)  . Each ouHeterogeneity is on average lowest in meta-analyses assessing an obstetric outcome, with evidence of a difference compared with the largest group of meta-analyses comparing general health–related outcomes; the estimated ratio of variances is 0.39 . We find that heterogeneity is higher in meta-analyses examining resource use or hospital stay/processes and internal and external structure-related outcomes than those assessing general health–related outcomes. Similarly, heterogeneity appears higher in meta-analyses with various subjectively measured outcomes including consumption, satisfaction with care, composite end point , and surgical or device-related success/failure, compared with the reference group of meta-analyses assessing general health–related outcomes.About the types of intervention comparison, studies within meta-analyses evaluating a nonpharmacologic intervention are on average most heterogeneous. We find that heterogeneity is lowest in meta-analyses comparing two pharmacologic interventions.The estimated ratios of between-study variances in 3.3τ2) is t, which has a median of 0.03 and 95% range of 0.0002–5.16 on the untransformed scale.Initially, we report a predictive distribution for a future meta-analysis for a general setting. This was obtained from a Bayesian hierarchical model fitted to all meta-analyses in the data set, including no meta-analysis characteristics as covariates. The fitted distribution for log(t distributions for log(www.jclinepi.com (Section A.3). Although the inverse-gamma distribution does not provide the best fit for underlying values of between-study variance in a meta-analysis, we provide predictive inverse-gamma distributions for www.jclinepi.com (Section A.3). These distributions would facilitate Bayesian random-effects meta-analysis with a conjugate prior for the between-study heterogeneity variance. In Bayesian analysis, use of a conjugate prior is sometimes preferred because the resulting posterior distribution is of the same known form as the prior.t distributions for between-study heterogeneity in two example settings. For a pharmacologic vs. placebo/control meta-analysis measuring an obstetric outcome, the predictive distribution gives little support to values above 1, whereas the predictive distribution for a nonpharmacologic meta-analysis measuring resource use gives moderate support to values of τ2 up to 10. Additional density plots representing predictive t distributions for between-study heterogeneity in a variety of settings are displayed in the www.jclinepi.com (Section A.3).The discrepancies among these fitted distributions reflect the comparisons of between-study variances in 3.4τ2 in a continuous-outcome meta-analysis, we reanalyzed data from two published meta-analyses. Both example meta-analyses represent the typical situation in which there are only a small number of studies in the meta-analysis, and Bayesian estimation is particularly beneficial. The first example meta-analysis consists of just four studies to compare exercise vs. control (no exercise or placebo exercise) with respect to depression in adults with chronic kidney disease , I2 = 79%]. The confidence interval for the conventional estimate of τ2 was obtained iteratively via the Q-profile method To demonstrate the use of an informative prior for the between-study variance  disease A [12]. Iτ, we used a uniform  prior, as recommended by Spiegelhalter et al. τ, which has been used as a prior in earlier applications to meta-analysis t prior for τ2 leads to a reduced estimate for the between-study heterogeneity of 0.19 . This approach incorporates our beliefs about the likely extent of heterogeneity in this setting, and we therefore consider these results more credible than those obtained using alternative approaches.Results for performing Bayesian random-effects meta-analysis with noninformative priors for heterogeneity are provided in τ2. The simple code for fitting each of the Bayesian models using informative priors for the between-study variance is available in the www.jclinepi.com (Section A.4). Central estimates for the summary SMD are similar, irrespective of the form of the prior distribution for heterogeneity. We note that central estimates for the between-study variance are also quite comparable across results from performing Bayesian meta-analysis with log-t or inverse-gamma prior distributions. However, there are noticeable discrepancies between the 95% intervals for both the combined SMD and τ2.Also presented in τ2 = 0 , I2 = 0%]. Bayesian meta-analysis using an informative log t prior for τ2 leads to a slightly increased estimate for the between-study heterogeneity of 0.002 . Although the central estimate for τ2 is only a little higher than in the conventional meta-analysis, this approach leads to a wider interval for the summary SMD because it allows appropriately for the uncertainty in between-study heterogeneity.As a contrasting example, we also reanalyzed data from a published meta-analysis consisting of just five studies to compare budesonide at different doses for chronic asthma B [15]. It or inverse-gamma prior distribution for the between-study heterogeneity variance.For this example in which the conventional heterogeneity estimate is low, central estimates and intervals for the summary SMD and the between-study variance show strong similarity between Bayesian meta-analyses using a log-4In this work, we have analyzed data from 6,492 continuous-outcome meta-analyses to describe predictors of heterogeneity and construct informative prior distributions for the between-study variance. We have demonstrated how these priors can be implemented in a Bayesian meta-analysis and given examples in which the precision of heterogeneity is improved with their use.t and inverse-gamma prior distributions for τ2 in continuous-outcome meta-analyses. The inverse-gamma distributions would facilitate Bayesian meta-analysis with a conjugate prior for the between-study variance. In two example meta-analyses, Bayesian meta-analysis with an informative prior for heterogeneity led to more precise estimates for heterogeneity and results were similar, regardless of the distribution of the informative prior.The results of the present study are consistent with those of the earlier work published on binary outcome meta-analyses www.jclinepi.com, an analyst could implement the prior for a general setting. In cases where no relevant data-based prior is available, researchers could use elicited opinion from experts to construct an informative prior for heterogeneity among studies in the meta-analysis.An important limitation lies in the fact that there are insufficient data for meta-analyses with certain characteristics. Given the rather extreme levels of heterogeneity observed and the low frequencies of meta-analyses specializing in cancer and respiratory diseases for many settings, we would be cautious about using our informative prior distributions in future meta-analyses related to cancer or respiratory diseases. A well-established problem in conducting Bayesian meta-analysis is the sensitivity of results to priors for variance components An issue that has not been addressed in this article is that results for the influence of medical area are highly prone to confounding. All Cochrane reviews in the CDSR have been prepared by authors, under the supervision of a Cochrane Review Group (CRG) in the Cochrane Collaboration. Because CRGs focus on a specific topic area, differences observed between disease areas may be caused by CRG editorial policies. Our results show extremely high heterogeneity among meta-analyses for cancer in comparison with other medical areas. Further examination of these meta-analyses revealed that this high estimate could be due to meta-analyses included in a single Cochrane review τ2 are restricted to use in meta-analyses performed on the SMD scale. As a simple solution, we could transform the between-study heterogeneity variance based on the SMD scale to that based on the MD scale by multiplying by a “typical” within-study standard deviation. However, it is difficult to obtain a good estimate for a “typical” standard deviation of outcome among participants in a study. An alternative to mean difference measures would be to use relative effects by computing a ratio of mean (RoM) values. Although the RoM may be desirable for ease of interpretation and statistical properties I2 as a statistic to quantify the degree of inconsistency among results of included studies in a meta-analysis. This commonly reported measure of inconsistency directly relates to the between-study variance and has the same interpretation regardless of the scale on which meta-analysis is performed. Although it is convenient to assign a prior to τ2, where possible, because this parameter is used in the analysis, we plan to construct informative prior distributions for I2 for use in future meta-analyses using different scales. Empirical evidence on I2 would provide useful information about the degree to which we would expect inconsistency across studies to reduce, on average, if meta-analysis was performed on a different scale or using a different type of outcome data.A limitation of this work is that the reported informative prior distributions for In summary, between-study heterogeneity was found to be strongly influenced by the type of outcome measured in the meta-analysis. Informative priors for heterogeneity would be useful in meta-analyses including few studies. Taking into account the important influences of meta-analysis characteristics on heterogeneity, implementing an informative prior for the between-study variance in a new meta-analysis would be beneficial in many settings.


TNF-α-308G/A polymorphism and obstructive sleep apnea syndrome (OSAS).The aim of our study was to investigate the association between the TNF-α-308G/A polymorphism and risk of OSAS.The Medline, Web of Science, EMBASE, Chinese National Knowledge Infrastructure (CNKI), and Cochrane Central Register of Controlled Trials were searched. Pooled odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to study TNF-α-308G/A polymorphism was significantly associated with risk of OSAS . In the subgroup analysis by ethnicity, the statistical similar results were observed both in European  and Asian population . When stratified by age, a significantly increased risk was observed in adult carries A allele compared with G allele , whereas no association was found in children .10 case-control studies were included in our meta-analysis. The results from our study showed that the TNF-α- 308G/A polymorphism contributed to the susceptibility to the risk of OSAS. Additional well-designed large studies are needed to validate our findings.Our study suggested that the  Obstructive sleep apnea syndrome (OSAS) is a common sleep disorder characterized by repetitive partial or complete obstruction of the upper respiratory tract during sleep, resulting in apnea or hypopnea TNF-α gene may directly influence the transcription of the TNF gene. The TNF-α gene is located within the highly polymorphic major histocompatibility complex (MHC) region on the short arm of chromosome 6p21.3 TNF-α gene have been identified. Among which, polymorphism at position −308 in the promoter region, consisting of a guanine (G) to adenine (A) substitution, has been reported to be associated with increased production of TNF-α levels both in vitro and in vivoTNF-α-308G/A polymorphism is strongly associated with circulating TNF-α concentrations, it is assumed that it might be closely related to the OSAS risk. To investigate a possible association between TNF-α polymorphism and risk of OSAS, we conducted a meta-analysis from all available relevant studies.Tumor necrosis factor (TNF)-α, a member of the TNF/TNFR cytokine family, is an intercellular communicating molecule involved in a wide variety of human diseases. Krueger et al. The Medline, Web of Science, EMBASE, Chinese National Knowledge Infrastructure (CNKI) and Cochrane Central Register of Controlled Trials were searched. A broad search strategy was used for optimum sensitivity. Using Medical Subject Headings (MeSH) and text words, we adopted the following terms to search the databases: (“obstructive sleep apnea–hypopnea syndrome” OR “obstructive sleep apnea” OR “sleep apnea” OR “apnea” OR “OSAS” OR “OSA”) AND (“single nucleotide polymorphisms” OR “SNP” OR “polymorphism” OR “gene variant” OR “mutation”) AND (“tumor necrosis factor α” OR “tumor necrosis factor-α” OR “tumor necrosis factor” OR “TNF-α” OR “TNF”). The search was restricted to humans. Titles and abstracts were screened up to 31 March 2014 were retrieved. Articles were screened at the title and abstract phase by two authors (Yanping Wu and Chao Cao).TNF-α- 308G/A polymorphism and OSAS susceptibility; (b) case–control study; (c) validated genotyping methods were used; (d) detailed genotype frequencies in cases and controls for the calculation. Major reasons for exclusion of studies were: (a) no control group, case reports or review articles; (b) duplicate data; (c) enrolled patients were using TNF inhibitors.Inclusion criteria for studies included: (a) evaluation of the relationship between Information was carefully extracted from all eligible studies independently by two of the authors, according to the inclusion criteria listed above. From each study, data on general study characteristics was extracted: first author's surname, year of publication, ethnic descent of the study population , genotyping methods, numbers of eligible cases and controls, genotype distributions in cases and controls and available subgroups. Disagreements were resolved by discussion and consensus.The Newcastle-Ottawa Scale (NOS), which was developed to assess the quality of non-randomised studies, was performed to assess the quality of included studies TNF-α-308G/A polymorphism was estimated for each study using the odds ratio (OR) and 95% confidence interval (CI). Between-study heterogeneity was assessed with the χ2-based Q statistical test 2I to efficiently test for the heterogeneity, with 2I<25%, 25–75% and >75% to represent low, moderate and high degree of inconsistency, respectively P>0.1 or 2I<50%, indicating a lack of heterogeneity, and for these analyses, the fixed-effect model was used; otherwise, the random-effect model was applied P<0.05 was considered significant. Effects were tested in models as follows: allele comparison (A vs. G), homozygote comparison (GG vs. AA), heterozygote comparison (AG vs. AA), dominant model (AG+GG vs. AA), and recessive model (GG vs. AA+AG). Stratified analysis was performed by ethnicity and age. Sensitivity analysis was performed to evaluate the effect of each individual study on the pooled ORs. Funnel plots and Egger's linear regression test were used to evaluate the potential publication bias All analyses were performed on Review Manager version 5.0.16 and Stata version 12.0. The Hardy-Weinberg equilibrium (HWE) was utilized to compare the observed genotype frequencies with expected genotype frequencies in controls. The association between risk of OSAS and TNF-α-308G/A polymorphism and OSAS. By further full text screening, 2 studies were duplicate publications and 2 had no interested outcomes. Finally, 10 full-text articles, with a total of 1522 OSAS cases and 1234 controls, which definitely addressed the association between TNF-α-308G/A polymorphism and OSAS risk were enrolled in our study TNF-α-308G/A polymorphism, four were performed in European population The search strategy retrieved 88 potentially relevant studies. After initial search by the titles and abstracts, 74 studies were excluded. Among the excluded studies, 4 were reviews or meta-analysis and 70 were not relevant to TNFα-308G/A polymorphism with OSAS risk  and Asian population , but not for children .Our study suggested a significant association of the Sensitivity analysis was carried out to assess the influence of each individual study on the pooled ORs. By removing each study, the pooled ORs were not altered significantly, which indicated that no individual study significantly affected the pooled results. When one HWE-violating study was excluded, the corresponding pooled ORs were still meaningful, showing the results were robust .Begg's test and funnel plot were performed to verify the publication bias of the literature. The shapes of the funnel plots did not show obvious asymmetry . The asyTNFα-308G/A polymorphism and OSAS risk. 10 studies involving 1,522 cases and 1,234 controls met the eligibility criteria. Our results demonstrated that the TNF-α-308G/A polymorphism was associated with OSAS risk . The A allele, compared with the G allele, was more likely to promote the risk of OSAS.In this meta-analysis, we concluded the existing data on the association of TNFα-308G/A polymorphism and OSAS risk remained significant in Europeans  and Asians . Genetic background might have significant influence on the susceptibility of OSAS. However, up to date, no studies were performed in Africans, which limited our further analysis in those populations. At the same time, it was worth emphasizing that TNFα-308G/A polymorphism was contributed to the increase of OSAS susceptibility in adults , but not for children . This might reflect the fact that increased age modified the associations and might helpful to understand the pathogenesis of adults' OSAS.In stratified analyses, we observed that the association between TNFα-308G/A polymorphism associated with risk of OSAS. Compared with G allele, A allele was closely related to increased susceptibility to OSAS. The significance of this polymorphism could be explained by its possible influence on the expression of TNF-α protein. It was reported that the A allele of TNF-α gene increased TNF-α expression and resulted in higher TNF-α production by means of regulating adipose tissue metabolism α played an important role in regulating the pro-inflammatory cytokines during sleep and closely related to the level of high sensitivity C-reactive protein (hsCRP) The results from our study showed that TNF-α-308 polymorphism increased susceptibility to the OSAS in both in European and Asian population. In addition, we firstly observed TNF-α-308 polymorphism increased the risk of OSAS in Adults, but not for children. Thirdly, no significant between-study heterogeneity and publication bias were observed in our study. Moreover, we carried out sensitivity analyses to evaluate the effect of each individual study on the pooled ORs.Compared with the previous meta-analysis performed by Huang et al Some limitations of this meta-analysis should be addressed. First, the number of published studies, especially for African population and children, was not large enough. Second, one study did not agreed with HWE in our analysis TNF-α-308 polymorphism increased susceptibility to the OSAS. These results provided the important role of the TNF-α-308G/A polymorphisms in the development of OSAS. Additional well-designed large studies were needed to validate our findings.In summary, this meta-analysis suggested that the A allele of the Checklist S1(DOCX)Click here for additional data file.Table S1Quality assessment of case-control studies.(DOC)Click here for additional data file.


I2 = 70.6). Nine of the 10 studies came from Europe, and the association was significant between drinking water magnesium level and the risk of CHD mortality . In conclusion, drinking water magnesium level was significantly inversely associated with CHD mortality.Epidemiological studies have demonstrated inconsistent associations between drinking water magnesium levels and risk of mortality from coronary heart disease (CHD); thus, a meta-analysis was performed to assess the association between them. Relevant studies were searched by the databases of Cochrane, EMBASE, PubMed and Web of Knowledge. Pooled relative risks (RR) with their 95% CI were calculated to assess this association using a random-effects model. Finally, nine articles with 10 studies involving 77,821 CHD cases were used in this study. Our results revealed an inverse association between drinking water magnesium level and CHD mortality (RR = 0.89, 95% CI = 0.79–0.99,  Coronary heart disease (CHD) killed almost seven million people worldwide in 2010, accounting for the largest fraction of death causes and years of life lost .It has previously been reported that drinking magnesium-rich water may be beneficial . A priorThe relevant articles were identified by searching the databases of Cochrane, EMBASE, PubMed and Web of Knowledge written in English. The following search strategy was carried out: (magnesium OR Mg OR drinking water) AND (coronary heart disease (CHD) OR myocardial infarction (MI) OR ischemic heart disease (IHD)) AND . The references list was also reviewed to include all the related articles.The current analysis included all the observational studies which reported drinking water magnesium levels and CHD mortality risk. The outcome measure was the incidence of the mortality of CHD or IHD or MI. The exposure of interest was drinking water magnesium levels. All included studies provided the relative risks (RR) or odds ratio (OR) and their 95% confidence intervals (CI), or provided enough data to calculate them. If the articles werefrom the same study or same populations, we then included only the most recent study; reviews or letters to the editor were excluded.All papers that were chosen based on the detailed search strategy were read by two of the authors (Lei Jiang and Pengcheng He). The following information was taken from each study: the publication years, study type, geographic locations, sex, number of participants, the results of CHD mortality outcome and the adjustment factors. The RR estimates and 95% CI for drinking water magnesium levels and CHD mortality risk were also extracted. We extracted the multivariable RR and its 95% CI if possible. Otherwise, we abstracted the crude RR estimates.2I statistics [p value of <0.05 represented significance.The inverse variance-weighted mean of the logarithm of RR and their 95% CI was calculated to evaluate the association between drinking water magnesium levels and CHD mortality. A random-effects model was used to combine study-specific RR (95% CI) . Heterogatistics . The smaatistics . Sensitiatistics  was perfFive hundred eighty-seven articles from PubMed, 432 articles from Cochrane database, 561 articles from EMBASE and 632 articles from the Web of Knowledge were selected by our search strategy. After reviewing the title/abstract, 72 articles were reviewed in full, 63 of which were subsequently excluded for various reasons. One paper reported the males and females respectively for the association between drinking water magnesium levels and the risk of CHD mortality . TherefoI2 = 70.6%)  .I2 = 70.6%, Pheterogeneity = 0.000) was found. Thus, we conducted univariate meta-regression to explore the cause of this high heterogeneity with the covariates of publication year, location, sex, study type, CHD mortality outcome, and number of cases. No significant association was found in the above covariates.In the pooled results, high heterogeneity . However, the association was nonsignificant in the cohort studies. In subgroup analyses for geographic locations, the highest levels of magnesium in drinking water versus the lowest levels were significantly associated with reduced risk of CHD mortality in Europe . There were three studies that reported the association in males and two studies in females. However, no significant associations were found either in males or females. Furthermore, in stratified analysis by CHD mortality outcomes, the association was only significant in the MI group . Detailed results for the overall and subgroup analyses are shown in p = 0.475).Sensitivity analysis  did not We conducted the first comprehensive meta-analysis that looked at an association between drinking water magnesium levels and CHD mortality risk. The findings from this meta-analysis suggest that the high levels of magnesium in drinking water may reduce the risk of CHD mortality. The association was also found in case-control studies, as well as in the European population.In our pooled results, we found evidence of high between-study heterogeneity. A previous study  had repoA highlight of this study is the large sample size  we were able to utilize in finding an inverse association between drinking water magnesium levels and CHD mortality risk. But, there are also some limitations in our study. First, nine of the 10 studies were conducted in Europe, and only one study was in China. A significant association was found only when we pooled the results for the subgroup of European studies. Notably, six of the ten studies came from Scandinavia (Sweden and Finland). Therefore, the results may be more applicable to the European population, and especially to Scandinavian populations. More studies originating in other countries are required to assess this association. Second, our study included seven retrospective studies and three cohort studies. For retrospective studies, original studies could cause some recall or selection bias, but proper epidemiological method requires clarification of the association in the original article. In our results, the association was only significant in retrospective studies, but not in cohort studies. This is because only three studies included in this study were cohort design. Therefore, further studies with cohort design are required. Third, there are only few studies which report the association between the level of magnesium in drinking water and CHD mortality for males and females respectively. Although we pooled the results for the subgroup analysis by sex, the small sample size of studies lead to lower statistical power. Finally, we found evidence of high between-study heterogeneity in the pooled analysis and some subgroup analyses. However, meta-regression could not explain this high heterogeneity. Thus, some other genetic and environment variables may affect this high heterogeneity.In conclusion, findings from this study showed that a higher level of magnesium in drinking water was significantly inversely associated with the risk of CHD mortality, especially among Scandinavian populations.


Adenomatous polyposis coli (APC) gene polymorphisms are believed to contribute to tumor susceptibility. However, the association between genetic variants (A/T) in the APC gene D1822V polymorphism and colorectal cancer (CRC) susceptibility remains unknown. To determine this association, a case-control study was performed. The genotype of the APC gene D1822V variants was analyzed by DNA sequencing in blood samples collected from 196 patients with CRC and 279 healthy subjects. There were no significant associations between the case and control groups in the distribution of AT [odds ratio (OR), 0.604; 95% confidence interval (CI), 0.355–1.029) and TT genotypes  relative to the AA genotype. The ratio of the T allele was significantly lower (P=0.047) in the case group compared with the control group , indicating that the T allele conferred a protective effect in CRC. The frequency of the AT genotype among the subjects diagnosed at >45 years of age was lower than those diagnosed at a younger age (P<0.05). The present study demonstrates that the T allele of the D1822V polymorphism may exert a protective effect against CRC, however, these findings require further validation in a larger sample size. Colorectal cancer (CRC) is the third most common malignancy in the world and is one of the most frequent malignancies in China . In ShanAdenomatous polyposis coli (APC) has been identified as a tumor suppressor gene, located on chromosome 5q21-22, with a full-length of 11,025 bp. The cDNA sequence of APC is composed of 8,538 nucleotides that encode 2,843 amino acids .Gene polymorphisms define the predisposition to human disease that arises from variations in genome sequences ,5. PolymMutations in the APC gene have been shown to be responsible for the autosomal dominant inherited disease, familial adenomatous polyposis (FAP). The types of mutation include missense mutations, small insertions or deletions. These mutations result in the early appearance of the stop codon, with a subsequent generation of truncated APC protein , and abeThese findings indicate that it is possible that a single nucleotide polymorphism (SNP) may form the genetic basis for different individuals with FAP, with similar lifestyles to each other, having varied susceptibilities to CRC. D1822V is the most common APC variant described in the literature . HoweverThe present case-control study consisted of 196 patients with newly diagnosed and histopathologically confirmed primary CRC who had not previously undergone radiotherapy or chemotherapy. The patients were all admitted to the Zhong Nan Hospital of Wuhan University  between February 2010 and November 2011. The control group included 279 histopathologically-confirmed cancer-free subjects with no current or previous diagnosis of cancer, who underwent health examinations at the Zhong Nan Hospital during the same period. The cases and controls were all residents of Hubei, and were matched for age and gender frequency using a uniform questionnaire, which included demographic characteristics, details of medical and family history and other information.The study was approved by the Investigation and Ethics Committee of the Zhong Nan Hospital and was undertaken in accordance with the Declaration of Helsinki 1975. All subjects provided written informed consent prior to participating.Venous blood (20 ml) was obtained in the morning from all the cases. Genomic DNA was isolated from whole blood using a Blood Genome DNA Extraction kit (Tiangen Biotech (Beijing) Co., Ltd., Beijing, China). Ultraviolet spectrophotometry was used to detect the content and concentration (A260/A280>1.8) of total DNA. Agarose gel electrophoresis  was used to detect DNA integrity. The DNA samples were diluted to 100 ng/ml in sterile triple-distilled water and stored at 4°C.2O. The PCR profile consisted of an initial 5-min denaturation step at 94°C, followed by 32 cycles of 30 sec at 94°C, 30 sec at 58°C and extension for 45 sec at 72°C, and a final 10-min elongation step at 72°C. The final PCR product was 410 bp.The upstream, 5′-ACCCAACAAAAATCAGTTAGATG-3′, and downstream, 5′-GTGGCTGGTAACTTTAGCCTC-3′, primer were provided by Invitrogen Life Technologies . PCR reactions were performed using a 20-μl reaction system containing 2 μl DNA template (100 ng/μl), 0.4 μl of each primer, 2 μl dNTPs, 0.2 μl ExTaq enzyme, 2 μl 10× PCR Buffer and 13 μl ddHAgarose gel (1%) electrophoresis  was used to recover the PCR products. The PCR fragments were sequenced at the Magic Biotech Company .2 tests were used to analyze demographic variables, environmental risk factors and APC genotype distributions between the case and control groups. Hardy-Weinberg equilibrium for genotypic frequencies was determined using the goodness-of-fit χ2 test. Univariate and multivariate unconditional logistic regression analyses were used to obtain odds ratios (OR) and corresponding 95% confidence intervals (CI). Stratified analysis was used to assess the effect of various factors on the incidence of CRC. P<0.05 was considered to indicate a statistically significant difference.Data analysis was performed using SPSS version 17.0 software . Bilateral χThe direct sequencing histogram of the PCR products is shown in The characteristics of the cases and controls are shown in The APC gene AA genotype was present in 226 subjects in the normal control group, the AT genotype was found in 50 subjects and the TT genotype in three subjects. The T allele frequency was 0.1004 and the A allele frequency was 0.8996. The genotype distributions of APC D1822V were in agreement with the Hardy-Weinberg equilibrium (P=0.95).In the case group, the AA genotype was present in 172 cases, the AT genotype in 23 cases and the AT genotype in one case. The T allele frequency was 0.0681 and the A allele frequency was 0.9362. The results are shown in There were no significant differences between the case and control groups with respect to the ratio of the AT  and TT  genotypes compared with the AA genotype . The proThere was no significant association between the APC D1822V SNP and the risk of CRC among subjects of a different gender or smoking and alcohol status. In addition, there was no link between the CRC risk and pre-existing hypertension or diabetes . HoweverThe effect of the D1822V SNP in APC was further evaluated to investigate possible associations with the clinicopathological characteristics of CRC. However, the variant homozygote TT genotype showed no significant association with the depth of tumor invasion, lymph node metastasis or the tumor location .The development of CRC has been shown to result from an interaction of multiple factors. APC gene mutation is a cause of the inherited autosomal dominant disease, FAP. The types of mutation involved in this disease process include missense mutations, small insertions or deletions. Each of these has the potential to advance the stop codon, resulting in the generation of a truncated APC protein, with altered function . MismatcAssociations between genotype and phenotype indicate that different types of APC mutations will have varying clinical manifestations . AlthougCodon 1,822 is located between the 4th and 57th amino acid repeat region in the APC protein. The wild-type in this area has the effect of downregulating β-catenin . The speRecently, a review and meta-analysis indicated that APC is a candidate gene for colorectal neoplasia susceptibility, and provided evidence that APC mutation may result in aberrant cellular proliferation, leading to the early stages of CRC . Based oThere are currently >20 ways to detect SNPs. These include TaqMan, mass spectrometry, chip methods and direct sequencing . Of thesThe majority of studies on SNPs use peripheral blood, which is more convenient and more economical, particularly for patients with unresectable lesions. A number of studies have confirmed that oncogene mutations, consistent with the primary tumors, are present in DNA extracted from peripheral blood –17. BaseIn another study, APC D1822V was shown to exert a protective effect in patients with colon  and rectal cancer , respectively . HoweverAPC D1822V is traditionally considered to be a risk factor for colorectal tumorigenesis. In 1999 , a VV hoAn increasing body of research has focused on investigating the association between the APC D1822V variant and CRC risk. In a previous study, the D1822V variant of the APC gene was found not to be associated with a family history of CRC, but it has the potential to increase the risk of adenoma transformation (Another large study  found anEach of these previously reported studies supports the findings of the present study, showing that the T allele for the APC gene D1822V variant appears to confer a protective effect against CRC. Other studies have shown that the APC Asp1822Val and Gly2502Ser polymorphisms are not associated with an increased risk of CRC or adenoma . HoweverThe present study lacks information on diet, and was therefore unable to analyze the interaction between diet, lifestyle and genetic factors. The study population was also relatively small, and larger studies will therefore be required to validate its findings. The present study does, however, provide evidence for the first time that the T allele of the APC gene D1822V polymorphism may be a protective factor for CRC in Chinese subjects, particularly in those >45 years of age.


P = 1.33E-35, OR = 1.23, 95% CI 1.20-1.27). Collectively, our analysis further supports previous findings that the rs16892766 polymorphism is significantly associated with CRC susceptibility. We believe that our findings will be very useful for future genetic studies on CRC.Colorectal cancer (CRC) is a common complex disease caused by the combination of genetic variants and environmental factors. Genome-wide association studies (GWAS) have been performed and reported some novel CRC susceptibility variants. The rs16892766 (8q23.3) polymorphism was first identified to be significantly associated with CRC in European ancestry. The following studies investigated this association in Chinese, Japanese, Romanian, Swedish, African American, European American, and Croatian populations. These studies reported consistent and inconsistent results. Here, we reevaluated this association using the relatively large-scale samples from 13 studies  using a meta-analysis by searching the PubMed, Google Scholar and CRCgene databases. We observed no significant heterogeneity among the included studies. Our results showed significant association between rs16892766 polymorphism and CRC ( Colorectal cancer (CRC), also called colon cancer or large bowel cancer, is the third most common form of cancer and the second leading cause of cancer-related death in the Western world and its lifetime risk in the United States is about 7%2345P = 3.30E-18, the odds ratio (OR) = 1.25, 95% confidence interval (CI) 1.19-1.32, Minor allele = C)6789101112131215P < 0.05)68913P > = 0.05)71011The rs16892766 (8q23.3) polymorphism was first identified to be significantly associated with CRC in European ancestry  to query the articles citing the studies and all references in these studies identified by the PubMed. We selected only published articles written in English. Theodoratou et al. report the first comprehensive field synopsis and creation of a parallel publicly available and regularly updated database (CRCgene) that catalogs all genetic association studies on colorectal cancer (http://www.chs.med.ed.ac.uk/CRCgene/)We searched the PubMed database to select all possible studies with key words including ‘rs16892766' and ‘colorectal cancer' or ‘8q23.3' and ‘colorectal cancer'. The literature search was updated on June 5, 2014. Meanwhile, we used the Google Scholar  the study must provided sufficient data to calculate the OR and 95% CI;We selected the studies meeting the following criteria: (1) the study was conducted by a case-control design; (2) the study evaluated the association between rs16892766 polymorphism and CRC; (3) the study provided the numbers of rs16892766 genotypes or (4) the study must provided sufficient data to calculate the numbers of rs16892766 genotypes or (5) the study provided an OR with 95% CI as well as the http://www.r-project.org/).We extracted the following information from each study: (1) the name of the first author; (2) the year of publication; (3) the population and ethnicity; (4) the numbers of AD cases and controls; (5) the genotype numbers of rs16892766 polymorphism in cases and controls; (6) the numbers of rs16892766 genotypes or (7) to calculate the numbers of rs16892766 genotypes; (8) the OR with 95% CI or (9) to calculate the OR and 95% CI; All relevant calculations were completed using the program R . I2 is a measure of heterogeneity and a statistic that indicates the percentage of variance in a meta-analysis that is attributable to study heterogeneityP < 0.01 and I2 > 50%.We evaluated the genetic heterogeneity among the studies included using Cochran's Q test, which approximately follows a http://cran.r-project.org/web/packages/meta/index.html).If there is no significant heterogeneity among the included studies, the pooled OR is calculated by the fixed effect model (Mantel-Haenszel), otherwise the OR is calculated by random-effect model (DerSimonian-Laird). Z test is used to determine the significance of OR. All statistical tests for heterogeneity and meta-analysis were computed using R Package .We evaluated the genetic heterogeneity of rs16892766 polymorphism among the selected studies using additive model and P = 1.33E-35, OR = 1.23, 95% CI 1.20-1.27). In As described above, we identified no significant heterogeneity among these 13 studies. We then performed a meta-analysis. We calculated the overall OR by the fixed effect model. Our results showed significant association between rs16892766 polymorphism and CRC using additive model .By excluding any one study, we identified that the association between rs16892766 polymorphism and CRC did not vary substantially. The funnel plots are symmetrical inverted funnels for models , which sP = 0.8239 and I2 = 0%). We then conducted a meta-analysis using fixed effect model. Our results showed significant association between rs16892766 polymorphism and CRC using additive model . Collectively, our analysis further supports previous findings that the rs16892766 polymorphism is significantly associated with CRC susceptibility. We believe that our findings will be very useful for future genetic studies on CRC.Recent GWAS identified rs16892766 (8q23.3) polymorphism to be significantly associated with CRC in European ancestryhttp://www.chs.med.ed.ac.uk/CRCgene/)Before our submission, we accessed the PubMed and Google Scholar databases using the key words ‘rs16892766' and ‘meta'. We identified two articles8P = 1.33E-35, OR = 1.23, 95% CI 1.20-1.27), which is more significant than previous GWAS Our study is different from previous studies8Pittman et al. generated a fine scale map of a 300 Kb region encompassing the rs16892766 association signal using 1,964 CRC cases and 2,081 controls6Despite these interesting results, our study has a limitation. Here, we investigated the association between rs16892766 and CRC using additive model. It is reported that most meta-analyses used an additive genetic modelG.Y.L., Y.S.J. and M.Z.L. conceived and initiated the project, searched the PubMed database and extracted the information from each study. G.Y.L., B.K.Q., X.S.Q., Z.H.Y. and G.Y.W. analyzed the data. R.N.F. and L.C.Z. prepared the figures 1–3. G.Y.L., Y.S.J., M.Z.L. and Y.Q.Z. wrote the manuscript. All authors reviewed the manuscript, and contributed to the final manuscript.


CYP1B1 , CYP3A4 , CYP3A5 , GSTM1/GSTT1 deletions, and SULT1A1 . Logistic regression was used to calculate the odds ratios (OR) with 95% confidence intervals (CIs), and unconditional logistic regression was used to estimate adjusted odds ratios (aORs) by ethnicity. Because of multiple testing, p values < 0.01 were significant after Bonferroni correction. SULT1A1 (c.638G>A) was associated to infant acute lymphoblastic leukemia and acute myeloid leukemia (AML) risk in males . CYP1B1 polymorphism was associated with a decreased risk of AML either for non-white or female children . Since polymorphisms of Cytochrome P450 genes presented gender-specific risk associations, we also investigated their expression. CYP1B1 was not expressed in 57.1% of EAL cases, and its expression varied by genotype, gender, and leukemia subtype. Maternal-fetal GSTT1 null genotype was associated with risk of EAL. This study shows that polymorphisms in genes of estrogen metabolism confer genetic susceptibility to EAL, mainly in males, and maternal susceptibility genes modify the risk for developing EAL in newborns.Based on observational studies, early age leukemia (EAL) was associated with maternal hormone exposure during pregnancy. We studied the association between genetic polymorphisms of estrogen metabolism and EAL. Using data from the Brazilian Collaborative Study Group of Infant Acute Leukemia (2000–2012), 350 cases and 404 age-matched controls and 134 mothers of cases and controls were genotyped to explore polymorphisms in genes of the estrogen metabolism pathway:  The acute leukemia causation pathway has been postulated to be a multistep process comprising an interaction between environmental  and genetic susceptibility factors . TogetheMLL, a master gene of hematopoiesis that regulates important genes of myeloid and lymphoid differentiation. Mutations arise in utero, as confirmed by backtracking methods using neonatal blood spots from children with acute leukemia [MLL rearrangements (MLL-r) resemble those found in secondary acute myeloid leukemia (AML) resulting from exposure to topoisomerase II inhibitors, what makes it a model for leukemogenesis in EAL with MLL-r. Thus, it is hypothesized that inadvertent exposure to biochemically similar topoisomerase II inhibitors during pregnancy may be responsible for DNA breaks in fetal cells and the underlying mechanism of EAL [MLL-r cases . The biological explanation for this observation was tested in a lymphoblastoid cell line, and a dose-dependent exposure to both estradiol and 4-hydroxiestradiol induced DNA double-strand breaks and MLL-r [Early age leukemia (EAL) has a short timeframe of exposure and the dose-effect in the causation pathways of EAL might be influenced by genetic susceptibility determined by multiple low-risk genetic variants. Most cases of EAL have somatic mutations, mainly rearrangements involving leukemia . MLL ream of EAL , 5. In om of EAL . The mosnd MLL-r . The ratCYP1B1  leads to a leucine to valine change at the protein level (p.L432V); the polymorphic allozyme exhibits greater activity toward estradiol hydroxylation than the wild-type [CYP3A4 generates an adenine to guanine substitution , although the phenotypic consequence is controversy and unknown [CYP3A5 íntron 3  leads to the formation of an aberrant splice site and to the absence of the enzyme when a guanine is located at the polymorphic site [GSTT1 and GSTM1 at 22q11.2 and 1p13.3, respectively. Both genes can be completely deleted (null genotype), leading to the absence of these proteins [SULT1A1 have been reported to decrease both the activity and thermostability of the enzyme  [Important features of cancer susceptibility genes are their functional heterogeneity, as genetic polymorphisms might be associated to different enzyme activity/levels. Therefore, several susceptibility genes involved in the same metabolic pathway would influence the risk of developing EAL. It is important to highlight that susceptibility of cancer in early-life may be quite distinct from that observed in adulthood cancer development. Both estrogen metabolism  and xenoild-type . RegardiCYP1B1, CYP3A5, GSTM1, GSTT1, and SULT1A1 in order to investigate whether these genetic changes modify the risk of developing EAL.Since previous studies have reported an association between estrogen intake and leukemia, it is likely that estrogen metabolic pathway generates carcinogenic compounds and genomic alterations in a precursor clone. Estrogen metabolism comprises several enzymes coded by polymorphic genes, thus allowing a considerable metabolic range in the population. Based on the hypothesis that genetic polymorphisms of estrogen metabolism modulates the amount of carcinogenic compounds a precursor cell may be exposed, we evaluated five polymorphisms in n = 350) diagnosed with either acute lymphoblastic leukemia  or AML (n = 115) were included. All controls (n = 404) were healthy and from the same Brazilian regions as the investigated cases. In addition, mothers of cases and controls who agreed to participate in this study by donating their biological material (n = 134) were genotyped.This case-control study included 754 children aged less than 24 months. Children were enrolled in the study from 2000 to 2012, and some cases and controls were included in previous reports that addressed questions related to maternal exposure during pregnancy . Patientwww.ibge.gov.br) [Race/ethnicity was defined according to Instituto Brasileiro de Geografia e Estatística, which gathers subjects based on self-defined ethnicity. The subjects were categorized as white or non-white. The non-white group consisted of the participants who self-identified as brown or black (.gov.br) .The exclusion criteria for both cases and controls were as follows: syndromes , T-cell precursor ALL, insufficient samples or bad DNA quality, and a lack of written consent from the mother.Leukemia was diagnosed by immunomolecular analysis of the lymphoid and myeloid cells according to standard criteria as described previously .All collaborating Brazilian institutions approved the study, and written informed consent was obtained from the mother or relatives responsible for the enrolled children in accordance with the Declaration of Helsinki. The Ethics and Scientific Committees of Instituto Nacional de Câncer-INCA and the National Research Ethics Committee  approved this study.Genomic DNA was isolated from peripheral blood cells or buccal cells . Among cases, DNA extraction was preferentially performed with samples lacking leukemic blast cells in order to rescue a maximum amount of constitutive DNA . Isolation was carried out using either the QIAmp DNA Mini kit  or Oragene DNA technology  according to the manufacturers’ protocols.CYP1B1 , CYP3A4 , and CYP3A5  polymorphisms were genotyped by restriction fragment length polymorphism according to primer sequence designs described previously [CYP1B1, CYP3A4, and CYP3A5 amplification were endonuclease-digested with the restriction enzymes AleI, MboII, and SspI , respectively. The GSTM1 and GSTT1 homozygous deletion (null genotype) was detected with multiplex PCR in which an internal control (CYP1A1) was co-amplified [SULT1A1 non-synonymous variants  were genotyped by pyrosequencing on a PyroMark Q24  after PCR amplification. Primers  flanked the region containing both polymorphisms. All conditions for the assays are available on request.The eviously –18. The CYP1B1, CYP3A4, and CYP3A5 are described elsewhere [RNA was extracted from bone marrow samples using TRIzol reagent  according to the manufacturer’s instructions. Following DNase treatment , cDNA was synthesized using SuperScript III Reverse Transcriptase  with 3 μg of purified RNA as a template. Real-time PCR was performed using Platinum SYBR Green qPCR Super Mix on a Rotor-Gene Q real time cycler . PCR was performed using diluted cDNA template in a 15 μL reaction mixture containing 0.30 μM of each primer and 7.5 μL of SYBR Green qPCR Mix. A melting curve phase program was applied with continuous fluorescence measurement between 60°C and 95°C. The negative controls for each primer consisted of a reaction with no cDNA added. The primer sequences for lsewhere , 21.-ΔCt method [The mRNA levels were normalized to glyceraldehyde-3-phosphate dehydrogenase expression using the following primers: 5’-CAACAGCCTCAAGATCATCAGCAA-3’ (forward) and 5’-AGTGATGGCATGGACTGTGGTCAT-3’ (reverse). Relative expression was determined using the 2t method . The val2) tests. The frequency of expected genetic polymorphism was calculated using the Hardy-Weinberg (HW) law based on the allele frequency in the control group. p Values < 0.05 were considered significant. The polymorphism frequency in children and mothers was compared to the control group using Chi-square (χ2) and Cochran Armitage Trend Tests. The disease risk was verified for the dominant, recessive and additive inheritance models; χ2 and Cochran Armitage Trend Test were used to calculate odds ratios (ORs) with 95% confidence intervals (CIs) and p values; unconditional logistic regression estimated adjusted odds ratios (aORs). Further Bonferroni’s correction for multiple testing was used; considering that five polymorphisms were analyzed, p values < 0.01 (0.05/5) indicated statistical significant results. Risk associations were calculated in the Statistical Product and Services Solutions statistical package, version 18.0 , while Cochran Armitage Trend Test to evaluate inheritance models was analyzed with R software (http://www.r-project.org/). Statistical analyses concerning gene expression data were performed using GraphPad Prism 5  software. The comparison among genotypes in relation to CYP1B1 and CYP3A5 expression was performed using the Kruskal-Wallis and Dunn's Multiple Comparison tests. The Mann Whitney test was used for all other analyses.Sociodemographic variables  were compared between cases and controls using the Mann-Whitney and Pearson's chi-square (χn = 350) and controls (n = 404) is provided in MLL-r was 45.9%.The distribution of demographic and laboratory variables between cases  and SULT1A1 (c.638G>A) variants (p < 0.05). These deviations from HWE raise concern when genotypes are mistyped or a biased control population is included, however given that our population is known to be heterogeneous and it was equally stratified, we assume that the results reflect an intrinsic feature of this study population. Nevertheless, all results concerning CYP3A4 c.-392A>G and SULT1A1 c.638G>A are shown in the supporting information and were not reported as significant in this manuscript.For all controls, the allele frequencies for CYP1B1, CYP3A4, CYP3A5, GSTT1, GSTM1, and SULT1A1 according to ethnicity distribution among ALL  and AML are shown in GSTT1 null genotype increased the risk of ALL in children 13–24 months-old . The presence of the GG genotype of CYP1B1 decreased the risk of developing AML  in non-white children.The genotype frequencies of MLL-r, only the ALL cases were sub-divided (infant ALL and ALL 13–24 months). The distributions of CYP1B1, CYP3A4, CYP3A5, GSTT1, GSTM1, and SULT1A1 genotype frequencies according to leukemia subtype, with controls adjusted by ethnicity, are shown in SULT1A1 (c.638G>A) only in infants , although the result was not statistically significant after Bonferroni correction . Among AML cases, decreased risk was associated with the GG genotype of CYP1B1 .Because infants (≤ 12 months of age) have been relevant for ALL and associated with different B-cell precursor subtypes with SULT1A1 c.638G>A was negatively associated with infant ALL for the dominant  and additive models . On the other hand, the same polymorphism was associated with an increased risk of AML  even after Bonferroni adjustment. In females, CYP1B1 variant  was associated with decreased risk of AML.The risk associations were also tested in ALL and AML according to gender: males and females , though the significance did not remain after multiple testing adjustment (Another aim of this study was to focus on the sum of low penetrance variants of the xenobiotic system (phase I and phase II) that can enhance the risk of EAL. The combinations of justment .SULT1A1 (c.638G>A) AA genotype reduced the risk of EAL . The combined mother/child genotype was assessed to evaluate the association with EAL risk . However, both results lost their significance with Bonferroni correction.Considering that differences in susceptibility, mainly during the fetal period, are dependent on the maternal system and placental metabolism, maternal genotypes were also tested, as shown in EAL risk . The GSTCYP1B1 polymorphism (c.1294C>G) was associated with a decreased risk of AML, which contradicts the initial hypothesis, we tested whether the genotypes of CYP1B1, CYP3A4, and CYP3A5 correlate with their own gene expression (CYP1B1 expression was found in a greater proportion of CYP1B1 wild-type genotype patients (44%) than CG and GG genotype carriers . Accordingly, the majority of the CG and GG genotype carriers with ALL (85%) or AML (67%) did not express the gene. No gene expression was found in any of the patients tested for CYP3A4 (n = 40). Also, no significant correlation was observed between CYP3A5 genotypes and gene expression.Because pression . We haveCYP1B1, CYP3A4, and CYP3A5 expression varied according to gender. CYP1B1 expression was detected in 29% of females and 30% of males, whereas CYP3A5 expression was detected in 14% of females and 16% of males. No association was found between gender and CYP3A5 expression. Regarding CYP1B1 and leukemia subtype, higher levels of gene expression were detected in girls with ALL  genetic polymorphism significantly modifies the risk in AML patients, whereas SULT1A1 polymorphisms (phase II metabolism) modify the risk in ALL patients.A hurdle in EAL is to understand the pathways that lead leukemogenesis toward the lymphoblastic or myeloid phenotype. For example, age-specific subtypes are well-defined in ALL cases, whereas the influence of age in AML cases is less remarkable, possibly reflecting different etiologies. Therefore, we analyzed the genetic polymorphisms by stratifying the cases according to diagnosis (ALL vs. AML). Overall, our data show that the CYP1B1 (c.1294C>G) presents a gain of activity compared to the wild-type isoform [CYP1B1 may increase EAL risk by inducing the formation of quinones and reactive oxygen species; however, our results show that the GG genotype is associated with a decreased risk of AML. Helmig et al. demonstrated that CYP1B1 polymorphic homozygotes have lower mRNA expression [CYP1B1 expression could compensate for the higher protein activity, possibly explaining the decreased risk observed in the present study. Our observations were in accordance with Helmig et al.’s study, as approximately 80% of children with CG/GG genotypes did not express the gene. Remarkably, the immaturity of the xenobiotic-metabolizing system was verified in this study, as the majority of the children did not express CYP1B1, CYP3A4, and CYP3A5.The G allele of  isoform , 22, thepression . In addipression . Thus, lSULT1A1 variants with cancer, and the results have been controversial [SULT1A1 (c.638G>A) was associated with a reduced risk of ALL. This decreased risk of ALL is not compatible with our hypothesis. Estrogen and its derivatives, which stimulate proliferation, are deactivated by sulfation because this conjugating reaction yields estrone-sulfate, a circulating form of estrogen that may be excreted [SULT1A1 polymorphism decreases enzymatic activity and thermostability, SULT1A1 polymorphism may increase the bioavailability of estrogens, leading to the formation of DNA adducts [SULT1A1 polymorphism.Several studies have investigated the association of oversial , 26. In excreted . Because adducts , 28. Not adducts ; therefoSULT1A1 (c.638G>A) in males and acute leukemia was found. Studies performed in animal models have revealed that female mice express higher levels of SULT1A1 than male mice [SULT1A1 expression and its polymorphic variant may influence the metabolism of compounds associated with the initiation of EAL.EAL is more frequent in males than females, but the reasons for this discrepancy are not yet understood. We observed some genetic polymorphism associations related to gender. An association of ale mice , 30. Theale mice . We hypoSULT1A1 (c.638G>A) homozygous genotype in mothers was associated with a decreased risk of EAL. Because sulfated hormones serve as circulating and stock forms, the reduced activity of the polymorphic SULT1A1 may decrease the bioavailability of estrogen to the fetal circulation thus protecting it from DNA damage [GSTT1 null genotype in both mother and child was associated with an increased risk of EAL. A recent study found that maternal haplotypes of GSTM3/GSTM4 and GSTP1 increase the risk of childhood ALL [The influence of maternal metabolism of estrogens and other compounds on fetal exposure levels is important. EAL is known to be initiated in utero and the primary interface between maternal and fetal circulation is the placenta, and the placental micro-environment is thought to be very important for acute leukemia predisposition. Many of the environmental exposures with a capacity to alter fetal development are likely to be modulated by maternal and fetal genetic factors. The presence of the A damage . The GSThood ALL . Future SULT1A1 and susceptibility to childhood leukemia. Fourth, gene expression results were used as complementary data for understanding the associations between genetic polymorphisms and EAL. Finally, the enrollment of children from different Brazilian regions enables broader interpretation of data than studies performed in single cities and/or institutions.This study has some limitations. First, the sample size is impacted by the heterogeneity of leukemia subtypes  that could have distinct pathogenesis. Second, maternal samples were not available for every case and control, and exposure data was restricted to a small number of cases and controls  that had biological material available for genotyping. It limited the analysis of an interaction between estrogen, other xenobiotic exposures and genetic polymorphism data. However, this study also has some strengths. First, a great number of infant leukemia, an extremely rare disease, was included; these numbers allowed us, for example, to find differences regarding gender and to evaluate genotype combinations. Second, the hypothesis is well supported by previous observational and experimental data on the association between estrogen exposure and leukemia development. Third, this report is the first to evaluate the association between genetic variability in CYP1B1, CYP3A5, GSTT1 and SULT1A1 variants exhibit significant risk associations with EAL. Gender is an important modulator of risk and may explain certain aspects related to the male/female incidence of EAL. Furthermore, genetic polymorphisms of mothers and their combination with child genotype supports the initial hypothesis and reveals that therisk association between EAL and polymorphisms of phase I and II estrogen metabolism are also dependent on both mother and child genetic susceptibility factors.In conclusion, S1 Table(DOC)Click here for additional data file.S2 Table(DOC)Click here for additional data file.S3 Table(DOC)Click here for additional data file.S4 Table(DOC)Click here for additional data file.S5 Table(DOC)Click here for additional data file.S6 Table(DOC)Click here for additional data file.S7 Table(DOC)Click here for additional data file.S8 Table(DOC)Click here for additional data file.S9 Table(DOC)Click here for additional data file.S10 Table(DOC)Click here for additional data file.


Hypoxia-inducible factor-2 alpha (HIF-2a) plays a major role in the progression of disease, although the role of HIF-2α gene polymorphisms in hepatitis B virus (HBV)-related diseases remains elusive. The aim of this study is to determine whether HIF-2a rs13419896 and rs6715787 single-nucleotide polymorphisms (SNPs) are associated with susceptibility to chronic hepatitis B (CHB), liver cirrhosis (LC), or hepatocellular carcinoma (HCC).A case-control study of 107 patients with CHB, 83 patients with LC, 234 patients with HCC, and 224 healthy control subjects was carried out, and the HIF-2a rs13419896 and rs6715787 SNPs were genotyped by polymerase chain reaction–restriction fragment length polymorphism (PCR-RFLP).p>0.05). However, in subgroup analysis by gender, the HIF-2a rs13419896 GA and AA genotypes were significantly associated with a risk of CHB  in females, and the A allele of rs13419896 was associated with a risk of CHB  and LC  in females. The rs6715787 CG genotype polymorphism may contribute to a reduced risk of LC in the Guangxi Zhuang Chinese population , as determined via subgroup analysis by ethnicity. Moreover, binary logistic regression analyses that were adjusted by drinking status indicated that the AA genotype of rs13419896 may contribute to an increased risk of LC in the non-alcohol-drinking population . In haplotype analysis, GG haplotype was significantly associated with a reduced risk of LC .No significant differences were observed in the genotype or allele frequency of two HIF-2a SNPs between the cases and controls (all The HIF-2a rs13419896 polymorphism is associated with an increased risk of CHB and LC in the Guangxi Chinese population, especially in females and in the non-alcohol-drinking population, while the HIF-2a gene rs6715787 polymorphism is associated with a decreased risk of LC in the Guangxi Zhuang population. Chronic hepatitis B virus (HBV) infection is a major public health problem throughout the world, particularly in sub-Saharan Africa and in most of Asia . ChronicTian et al. discovered HIF-2a in endothelial cells in 1997; it is also called endothelial PASprotein-1 (EPAS1), HIF-like factor (HLF), and HIF-related factor (HRF) . It is lGenetic polymorphisms are considered the main genetic elements involved in the development of various kinds of diseases . To dateA total of 648 subjects were recruited in this case-control study, including 107 patients with CHB, 83 patients with HBV-related LC, 234 patients with HBV-related HCC, and 224 healthy controls. All subjects were recruited at the First Affiliated Hospital of Guangxi Medical University from 1 June to 1 October 2014. The patients in the case group had tested positive for the hepatitis B surface antigen (HBsAg). Patients were included in the CHB group and LC group following the prevention guide of CHB diagnostic criteria , while they were included in the HCC group following the standard of hepatocellular carcinoma diagnosis and treatment criteria . To control for the effects of potential confounders, healthy participants were individually matched to the cases based on gender and age (±5 years).The patients infected with other hepatitis viruses, such as hepatitis C or hepatitis E, as well as those with liver diseases caused by non-viral infection, were excluded. None of the subjects had received treatment  before seeing a doctor.The study protocol was approved by the Ethics Committee of the First Affiliated Hospital, Guangxi Medical University , and wri5'- TTCCCTGTTCCCTCCTCCTTT-3' forward primer, 5'-TCCTACCCTGTGGTTGCCTCG-3' reverse primer of rs13419896 and 5'-TCAGTTGAGCATCCCTAACCC-3' forward primer, 5'-CCAGCCCTTGTCAGCATCTTT-3' reverse primer of rs6715787. Primer sequences and reaction conditions for the genotyping of HIF-2a polymorphisms are shown in All operations were carried out in strict accordance with the standard operating procedure. Genomic DNA was extracted from venous blood using the standard phenol-chloroform method. The genotyping of HIF-2a rs13419896 and rs6715787 SNPs was performed via polymerase chain reaction–restriction fragment length polymorphism (PCR-RFLP). Polymerase chain reaction (PCR) was used to amplify the target DNA, with the 2 test to verify the crowd representativeness of the study subjects. The χ2 test and Fisher exact test were used to analyze the distribution of the genotype and allele frequencies among the cases and controls. The independent risk and protective genotypes and alleles were selected by binary logistic regression analyses adjusting for age, gender, smoking status, drinking status, ethnicity, and body mass index (BMI). Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to assess the relative risk. SHEsis online software [http://analysis.bio-x.cn/myAnalysis.php).The distribution of the general demographic and clinical features between cases and controls was evaluated using the one-way analysis of variance test and the χ2 test for continuous and categorical variables, respectively. Hardy–Weinberg equilibrium (HWE) was employed with a goodness-of-fit χsoftware  was applp<0.05.All of the statistical analyses were performed using the Statistical Package for the Social Sciences . Statistical significance was assumed at two-sided values of A total of 648 participants (224 controls and 424 cases) were recruited in this case-control study. The demographic characteristics of the study are shown in p>0.05). To explore the impact of risk factors on CHB, LC, and HCC subjects, we conducted stratified analysis of confounding factors, such as gender, age, ethnicity, smoking, and alcohol drinking. Binary logistic regression analyses adjusted by gender are shown in p = 0.031 and OR = 12.506, 95% CI = 1.329–117.716, p = 0.027). Moreover, the A allele was also associated with a risk of CHB  and LC  in females. These results revealed that the GA and AA genotypes of rs13419896 may be risk factors for CHB in females in Guangxi, while the A allele genotype of rs13419896 may be a risk factor for CHB and LC in women in Guangxi. For the rs6715787 polymorphism, the CG genotype may contribute to a reduced risk of LC in the Guangxi Zhuang population , as shown in p = 0.034)  . No signp = 0.005).As the haplotype-based analysis is considered to have more power than SNP genotyping does, SHEsis online software was applied for the haplotype analysis of HIF-2α gene polymorphism (rs13419896 and rs6715787). We calculated the linkage disequilibrium in both the case and the control for the two SNP sites, and the value of the lowest frequency threshold for haplotype analysis is 0.03. According to the results, a total of four haplotypes  were derived from the observed genotypes. The haplotype distribution in the CHB, LC, and HCC patients and healthy controls are shown in Considering that the HIF-2α rs13419896 and rs6715787 genotype frequencies might be distinct among different populations, the genotype and allele frequencies of the two polymorphisms in our control group were further compared with those populations from the Haplotype Map (HapMap) project. The data in HIF-2, an important transcription factor in the hypoxic condition, was recently recognized as controlling the activation state and key microbicidal functions of immune cells . A growiThis is the first case-control study to reveal whether HIF-2a rs13419896 and rs6715787 polymorphisms are associated with susceptibility to CHB, LC, or HCC in the Guangxi population. In the present study, we attempted to explore the association between the HIF-2a gene rs13419896 and rs6715787 polymorphisms and HBV-related liver disease risk in a Chinese population. As a result, we found no significant differences in the genotype distribution or allele frequency of two HIF-2a SNPs between cases and controls. However, some confounding factors, such as gender, age, smoking, drinking, ethnicity, and BMI, probably play a role in the development of HBV-related liver disease, so we stratified our population to analyze the effect of these factors. In subgroup analysis, we showed that the GA and AA genotypes and the A allele of HIF-2a rs13419896 polymorphisms were associated with an elevated risk of HBV-related liver disease in female and non-alcohol-drinking populations, respectively. For the rs6715787 polymorphism, the CG genotype may contribute to a reduced risk of LC in the Guangxi Zhuang population. No significant differences were found after adjusting for age and smoking by binary logistic regression analyses. Furthermore, haplotype analysis indicated that the GG haplotype might be a protective factor for LC in the Guangxi population. According to our latest studies , 45 on tWith regard to rs13419896, which is located within intron1 of the gene, one study by Putra et al. , focusinWith regard to rs6715787, unfortunately, no research has investigated the association of HIF-2a rs6715787 polymorphism with various diseases, except for a report on living high-exercise high-training low (HiHiLo). However, one study showed that exon2 of the HIF-2a gene is the key to the DNA binding heterodimer . Since rA haplotype is a set of closely linked genetic markers present on one chromosome; such markers tend to be inherited together more frequently than expected by chance and could better explain the observed associations compared with each polymorphism independently. Therefore, further haplotype analysis should be performed to explore the associations between the HIF-2a gene and the risk of HBV-related liver disease. Our result indicated that the GG haplotype might be a protective factor for LC in the Guangxi population; it may induce splice-site variants  or affecIL-4, IL-6, IL-17, IL-23, INF-γet al.), which have crucial roles in innate immunity, are also related to the risk of CHB, LC, and HCC in our previous studies [HIFs contribute to inflammatory functions in various other components of innate immunity, such as mast cells, dendritic cells, and epithelial cells , which p studies –60. On tOur study has several limitations. First, the positive results were shown only in subgroup analysis. The potential skewing of results due to the relatively small sample size should been considered when interpreting the results. Second, the current study selected only two rather than all SNPs of the HIF-2α gene. In addition, we did not measure the serum HIF-2α levels both in patients and controls; thus, the relationship between the expression of HIF-2α and HIF-2α gene polymorphisms was unclear. Considering HIF-2a SNP as a useful clinical prognostic marker of HBV-related liver disease patients, much more extensive studies on mechanisms at the cellular and molecular levels are needed in the future. Finally, the study sample was limited to the Guangxi population. Because the genotype distribution exhibits regional and ethnic differences, further studies involving other ethnic populations are necessary for systematical study.In conclusion, HIF-2a rs13419896 polymorphism is associated with an increased risk of CHB and LC in the Guangxi Chinese population, especially in females or non-alcohol drinkers; meanwhile, the HIF-2a gene rs6715787 polymorphism is associated with a decreased risk of LC in the Guangxi Zhuang population. In addition, the GG haplotype of the two SNPs might be a protective factor for LC in this population. An effective risk prediction model should be constructed to further clarify the connection between HIF-2a polymorphism and HBV-related liver disease.S1 File(PDF)Click here for additional data file.S2 File(XLS)Click here for additional data file.S3 File(PDF)Click here for additional data file.


An increasing number of studies have highlighted the potential link between EXO1 polymorphisms and cancer risk, although no consensus has yet been obtained. Thus, we aimed to obtain a thorough and current assessment of EXO1 polymorphisms and cancer susceptibility by performing a meta-analysis. A comprehensive literature retrieval was performed on PubMed, EMbase, Web of Science and Wanfang databases. The odds ratio (OR) and 95% confidence interval (CI) were applied to assess the results. Finally, 39 case-control studies of the nine EXO1 polymorphisms that involved 21,651 cases and 21,348 controls met our inclusion criteria. The pooled analysis indicated that the rs1047840 polymorphism conferred a significantly increased susceptibility to cancer in an allelic model. Similarly, the rs3754093, rs1776177, rs9350, rs10802996, rs1635498, rs1776148 and rs851797 polymorphisms were also associated with an increased susceptibility to cancer in an allelic model, respectively, while no significant association was identified for rs1635517 polymorphism. For the rs1047840 polymorphism, in an ethnicity subgroup analysis, a significantly increased susceptibility to cancer for Asians was identified in all the genetic models, and for Caucasians in an allelic model. Our findings provide the evidence that the rs1047840, rs9350, rs10802996, rs1635498, rs1776148, rs1776177, rs3754093 and rs851797 polymorphisms may act as risk factors for cancer. Currently, cancer is a primary cause of human death, which can be attributed to its high rates of morbidity and mortality in the United States and many other countriesThe Exonuclease 1 (EXO1) gene, which belongs to the RAD2 nuclease family, encodes a member of the mismatch repair (MMR) system that plays a critical role in maintaining genomic stabilityRecently, the associations between EXO1 genetic polymorphisms and susceptibility to various type of cancers had been widely investigated. An EXO1 polymorphism at codon 589 rs1047840) is a non-synonymous single nucleotide polymorphism (SNP) that has been associated with susceptibility to lung cancer (LC)56840 is a We performed a comprehensive literature search using the PubMed, Web of Science, EMbase and Wangfang databases  in which we applied the following search terms: (EXO1 OR exonuclease 1) AND  AND . We also manually retrieved reference lists from these enrolled publications, aiming to ensure that all eligible studies were included.The detailed inclusion criteria were as follows: 1) the study was a case-control study; 2) the study evaluated the association between EXO1 polymorphisms and cancer susceptibility; 3) the study comprised useful allele and genotype frequencies to estimate the crude ORs at 95% CIs. However, all meta-analyses, reviews, animal studies and case-only studies, as well as those duplicated previous publications, were definitely excluded. Studies deviated from Hardy Weinberg Equilibrium (HWE), studies that were not concerned with cancer susceptibility and abstracts with incomplete genetic data were also removed from this analysis. When a case group was investigated in more than one publication, the publication with the largest number of participants was selected.The following information was extracted from each study by two independent investigators (Meng Zhang and Duran Zhao): name of the first author, year of publication, ethnicity, genotyping methods, source of controls, cancer type, total number of cases and controls, and HWE. Any discrepancies between the two investigators’ selections were resolved by consensus. Regarding the sources of controls, all eligible case-control studies were defined as either population-based (PB) or hospital-based (HB).2-based Q-test to evaluate between-study heterogeneity within the studiesP-value was less than 0.1. If there was no significant heterogeneity, a fixed effect model (Der-Simonian Laird) was used to evaluate the point estimates and 95% CI; otherwise, a random effects model (Der-Simonian Laird) was usedZ-test was used to determine the significance of overall ORs. In addition, Bonferroni corrections were also performed to adjust the resultsSusceptibility to cancer related to EXO1 polymorphisms was calculated directly from data provided in the eligible studies. Crude OR corresponding to 95% CI was applied to evaluate the strength of association between EXO1 polymorphisms and cancer susceptibility. Variants in EXO1 included rs1635498, rs1047840, rs851797, rs3754093, rs1776177, rs1776148, rs1635517, rs10802996 and rs9350. Each variable was analysed in allelic comparison, heterozygote comparison, homozygote comparison, dominant and recessive models. For example, the pooled ORs of EXO1 rs9350 polymorphisms were calculated by allelic comparison (T vs. C), heterozygote comparison (TC vs. CC), homozygote comparison (TT vs. CC), dominant (TT + TC vs. CC) and recessive models (TT vs. TC + CC). We used a χhttp://hapmap.ncbi.nlm.nih.gov/cgi-perl/gbrowse/hapmap3r2_B36/) encompassing the polymorphisms in EXO1 evaluated in currently study2 statisticsWe extracted data from the 1000 genomes Project Phase III (P > 0.05) in controls population except for 11 case-control studies569101719As shown in 456891017181920212223For the rs1047840 polymorphism, 11 case-control studies with 6,289 cases and 6,333 controls met our inclusion criteria. Among these, four studies were of Caucasian individuals, five were of Asians and others were of mixed groups. For the rs9350 polymorphism, five case-control studies, which included 3,173 cases and 2,895 controls were enrolled. Among these, only one study was performed in a Caucasian cohort, while the others considered as Asian individuals. For rs1776177 polymorphism, three case-controls of Asians that encompassed 1,217 cases and 1,217 controls were enrolled. For the rs1776148 polymorphism, we ultimately enrolled two case-control studies of Caucasians that included 711 cases and 490 controls. For the rs10802996 polymorphism, two case-control studies of Asians that included 305 cases and 457 controls met our eligibility criteria. Moreover, for the rs851797, rs3754093, rs1635517 and rs1635498 polymorphisms, there were four case-control studies with 2,489 cases and 2,489 controls that were included in this present meta-analysis, and the ethnicity of these studies were of Asian populations.P = 0.000, P = 0.000, P = 0.000). Additionally, rs9350, rs10802996, rs1635498, rs1776148 and rs851797 polymorphisms were also conferred an increased overall susceptibility to cancer in an allelic model . However, no significant association was identified for rs1635517 polymorphism and overall cancer susceptibility.The results of rs1047840 and rs9350 polymorphisms and cancer risk were shown in P = 0.000). When the subgroup analysis was conducted by source of control, an increased susceptibility to cancer was identified for P-B groups in an allelic model . In addition, we uncovered an increased susceptibility to lung cancer (LC) in an allelic model . In the stratification analysis by genotyping methods, an increased susceptibility to cancer was revealed for PCR-RFLP  and SNP-Chip  groups, respectively.Results of subgroup analyses were presented in P = 0.000) for Asian population.For rs9350 polymorphism, when the stratification analysis was conducted based on genotyping method, we also identified a significant increased susceptibility to cancer in PCR-RFLP group in an allelic model . However, only six of them can be matched from the database, including rs1635498, rs1047840, rs9350, rs1635517, rs1776148 and rs1776177 polymorphisms. LD plots for TSI population presented high LD value (r2 = 0.70) between rs1635517 and rs1776177 polymorphisms, for CHB, JPT, CHD and GIH populations presented moderate LD values (r2 ≥ 0.50), while for YRI, ASW and LWK populations presented a lower LD value (r2 < 0.35) (2 < 0.35) between rs1776148 and rs9350 polymorphisms.In order to better understand these results, LD analysis was performed to test for the existence of bins in the region comprising these eight polymorphisms (Meta- < 0.35) . LD plotMultiple factors are involved in cancer formation and progression. Abundant evidence suggests that genetics play an important role in determining cancer susceptibility, and understanding associations between genetic polymorphisms and malignancies may provide personalized analysis and reveal the predictive value of certain carcinomas. An increasing number of research studies of tumourigenesis have highlighted EXO1 as a promising target.The DNA repair system is known to be essential for maintaining genetic stability and offering protection from cancer initiation. EXO1 is an exclusive exonuclease gene that participates in the human MMR system. Genetic disorders in the MMR system result in the absence of DNA MMR function, resulting in an increased frequency of spontaneous mutations, which may give rise to the steady accumulation of oncogenes and tumour suppressors, which eventually contribute to tumourigenesisth amino acid of the Exo1 protein from lysine to glutamic acid, which might affect EXO1 expression. For the rs1047840 polymorphism, it is located in an exonic splicing enhancer (ESE) regionFor rs1047840, one polymorphism in EXO1 is located on exon 12, and its variation leads to a change of the 589et al.P < 0.05). Similarly, many other molecular epidemiological studies are consistent with this association of the EXO1 rs1047840 with BC, LC, oral cancer (OC) and GC91023et al.et al.Pdiscovery = 6.6 × 10−4, Preplication = 0.039, Pjoint = 2.5 × 10−4; ORjoint = 0.80, 95% CI: 0.71–0.90), while carriers of the A allele had lower expression (P = 0.002).Owing to the importance of EXO1 polymorphisms, an increasing number of studies have explored the associations of EXO1 polymorphisms and cancer risk. Luo et al.et al.27et al.’s2 = 0.70) for TSI population, and for CHB, JPT, CHD and GIH populations presented moderate LD values (r2 ≥ 0.50), while for YRI, ASW and LWK populations presented a lower LD value (r2 < 0.35). However, our meta-analysis results seemed not to be in accordance with the LD analyses that rs1776177 polymorphism is a risk factor for cancer rather than rs1635517. Limited number of studies and sample size may account for this discrepancy, therefore, future studies are warranted to verify this finding.As these findings are not conclusive, we conducted the present comprehensive meta-analysis to shed light on the role of EXO1 polymorphisms in tumourigenesis. Unlike the previous work conducted by Duan etc., we are unable to validate the potential effects on these cancers homogeneous or not. In addition, as ten case-control studies were excluded from the pooled analysis, there retained only two case-control studies for rs10802996 and rs1776148 polymorphisms, respectively. On the basis of small samples, possibly leading to an underestimate of the true association. All these confusions should be further confirmed in a series of much larger studies. Secondly, the effects of EXO1 polymorphisms on cancer susceptibility might be affected by several factors, such as age, sex, smoking status and matching criteria. A lack of samples and data may cause inconsistency in results and lead to possible publication bias, which often affect the precision of overall results. Thirdly, tumours are polygenic diseases, and analyses of haplotype and gene-to-gene associations are also important because gene-to-gene and gene-to-environment interactions may modulate cancer susceptibility.There are some strengths and limitations in the present study. The most important strength is that we have conducted a comprehensive retrieval for all eligible studies and polymorphisms, and the sample size was markedly expanded, which helped to reveal some findings not suggested in the previous work. Secondly, studies deviated from HWE were excluded, ensuring the accuracy of the finally data. Simultaneously, several limitation should be discussed. Firstly, the phenotype of our study is a heterogeneous aggregation of a variety of cancer types, and only for rs1047840 polymorphism, a subgroup analysis based on cancer type was conducted, while for others, attributing to the limited number of studies for specific cancers, such as BC, CC and In conclusion, our meta-analysis suggests that the rs1047840, rs9350, rs10802996, rs1635498, rs1776148, rs1776177, rs3754093 and rs851797 polymorphisms in EXO1 may be risk factors for cancer. In the future, large-scale well-designed case-control studies are warranted to verify our findings.How to cite this article: Zhang, M. et al. Associations between Nine Polymorphisms in EXO1 and Cancer Susceptibility: A Systematic Review and Meta-Analysis of 39 Case-control Studies. Sci. Rep.6, 29270; doi: 10.1038/srep29270 (2016).