Due to the lack of strong evidence to identify the relationship between antihypertensive drugs use and the risk of prostate cancer, it was needed to do a systematic review to go into the subject.We systematically searched PubMed, Web of Science and Embase to identify studies published, through May 2015. Two evaluators independently reviewed and selected articles involving the subject. We used the Newcastle-Ottawa Scale (NOS) to assess the quality of the studies. All extracted results to evaluate the relationship between antihypertensive drugs usage and prostate cancer risk were pool-analysed using Stata 12.0 software.A total of 12 cohort and 9 case-control studies were ultimately included in our review. Most of the studies were evaluated to be of high quality. There was no significant relationship between angiotensin converting enzyme inhibitors (ACEI) usage and the risk of prostate cancer , according to the total pool-analysed. Use of angiotensin receptor blocker (ARB) was not associated with the risk of prostate cancer , while use of CCB may well increase prostate cancer risk based on the total pool-analysed . Moreover, subgroup analysis suggested that use of CCB clearly increased prostate cancer risk  in terms of case-control studies. There was also no significant relationship between use of diuretic  or antiadrenergic agents  and prostate cancer risk.There is no significant relationship between the use of antihypertensive drugs  and prostate cancer risk, but CCB may well increase prostate cancer risk, according to existing observational studies.The online version of this article (10.1186/s12894-018-0318-7) contains supplementary material, which is available to authorized users. The prevalence of hypertension is consistently high in older adults and regarded as a vital risk factor for cardiovascular diseases, congestive heart failure, and coronary heart disease , 2. AntiHowever, antihypertensive drugs and cancer risk have long been raised as concerns . It was The incidence of prostate cancer is increasing, and it is the main cause of cancer death in males in the Western countries . StudiesDue to there being the long-term follow-up in the observational studies, we performed a systematic review of observational studies to confirm whether the use of antihypertensive drugs was able to result in the prostate cancer in the human body.Our review was conducted in accordance with PRISMA  and MOOSE  guidelines.Studies were included if they met the following points: (1) Studies were cohort or case-control studies; and (2) the relationship between the use of one or more types of antihypertensive drugs and prostate cancer was reported in the study.We systematically searched PubMed, Web of Science and Embase to identify studies published through May 2015. The search terms were composed of the following: “beta blockers”, “angiotensin converting enzyme inhibitors”, “angiotensin receptor blockers”, “calcium channel blockers” “alpha blockers” “antihypertensive drugs” and “prostate cancer”. The details of the search methods are summarized in the Additional file Two authors independently evaluated the studies retrieved from the databases to select the studies that met the inclusion criteria. Disagreements between the two reviewers were resolved by discussion or in consultation with an arbitrator. The following information was extracted from the included articles by the two authors independently: study design, geographic region, type of medication, sex, age range, follow-up time, adjusted factors in each study (Table1). The RR (relative risk), HR (hazard ratio), OR (odds ratio), SIR (standardized incidence ratio) along with their corresponding 95%CI , all of which indicated the relationship between antihypertensive drugs and prostate cancer were abstracted. If there was missing information in the article, we contacted the authors via e-mail or telephone. We used data from a 2 × 2 table to recalculate crude estimates when the outcome measures were unsuitable for meta-analysis and we failed to gain the data from the authors.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). The scale provided eight items consisting of three subscales: selection of subjects (four items), comparability of subjects (one item) and assessment of outcome/exposure (three items). The highest scores were nine for the eight items because there were two scores in the comparability of subjects. A study with greater than or equal to seven scores was considered to be of high methodological quality.The quality of the observational studies was independently evaluated by two authors based on the Newcastle-Ottawa Assessment Scale(NOS) on the website (p < 0.1) and by estimating I2. If I2 was more than 40%, there was significant heterogeneity among studies [Extracted RRs, HRs, ORs and SIRs and their 95% CIs that were adjusted for most confounders were pooled to compute the RR between antihypertensive drugs and prostate cancer risk . The fou studies .Our search strategy yielded 729 records, A total of 193 and 491 records were excluded due to duplicated records and irrelevant subjects respectively. A total of 45 full-text articles were assessed and 21 studies that met our criteria were ultimately included . Subgroup analysis also showed no significant relationship between the use of ACE inhibitor and the risk of prostate cancer according to the poolanalysis of cohort studies  and case-control studies   . Subgroup analysis also indicated that without significant relationship between CCB usage and the risk of prostate cancer in terms of cohort studies  but there was clear association between the use of CCB and the risk of prostate cancer according to the pool-analysis of case-control studies    Fig. .Fig. 5FoThere were 5 and 4 studies involving the association between the use of diuretics , 36, 35 To the best of our knowledge, this is first meta-analysis to investigate the relationship between the use of antihypertensive drugs and the risk of prostate cancer. Our main findings suggested that there was no significant association between the use of ACE inhibitors, ARB, beta-blockers, diuretics or antiadrenergic drugs and the risk of prostate cancer. However, CCB usage appeared to be associated with the risk of prostate cancer. Moreover, that considerable heterogeneity existed among the studies resulted in the reduction of evidence levels.Prostate cancer is the most frequently diagnosed cancer diagnosed and the main causes of cancer death in older men in Western countries. Antihypertensive drugs such as ACE inhibitors and ARB are widely used for the management of hypertension and the prevention of cardiovascular disease events in high-risk persons , 40, 41.The topic of the relationship between the use of antihypertensive drugs and the risk of prostate cancer remain controversial, especially in the use of CCB and ACE inhibitors or ARB. The findings of a case-control study by Vezina, RM et al. suggested that there was no association between the use of CCB and the risk of prostate cancer in men younger than the age of 70 . Debes, ACEI and ARB have been successfully used as potent antihypertensive drugs for a long period of time, and some literatures have suggested that these drugs could serve as new anticancer drugs . The incA meta-analysis suggested that the use of BB was associated with reduced specific mortality among patients with prostate cancer . HoweverThe present systematic review and meta-analysis provided a summary analysis of previous relevant studies which could yield a conclusion characterized by compromise. According to the pooled analyses of cohort studies, we generally did not find significant associations between the use of antihypertensive drugs and the risk of prostate cancer. Nevertheless, CCB usage might contribute to the higher risk of prostate cancer, based on findings from the case-control studies. Moreover, antiadrenergic drugs had a protective effect on the development of prostate cancer, according to the meta-analysis of two cohort studies. As we know, there was longer duration of follow-up in observational studies and we only reviewed these studies to explore whether the long-term use of antihypertensive drugs affected the incidence of prostate cancer in natural population. In the present review, most of the studies had a follow-up time of at least five years, and we had a larger sample to analyse than any previous studies. The main confounding factors, such as age, BMI, and race were adjusted for in most of the studies. Although we included studies with samples containing females, the relative risk was calculated under an adjustment for sex. Moreover, most of the studies were assessed with a high quality of methodology. As a result, our pooled analyses provided a conclusion that was more closer to the truth.Our studies also had some limitations. First, our evidence grade was compromised by the considerable heterogeneity, which might be caused by various factors, including study design, race, follow-up time, social background. Second, small sample were analysed on the association between diuretics or antiadrenergic drugs and prostate risk, and the results should be cautiously considered. Third, we did not detect the publication bias, although it might exist among these studies. Finally, we did not conduct a dose-response meta-analysis due to the lack of relevant data in the included studies.We reviewed 21 observational studies and found that there was no significant association between use of antihypertensive drugs and the risk of prostate cancer. Nevertheless, CCB usage might contribute to the higher risk of prostate cancer based on findings from the case-control studies. Unfortunately, the high heterogeneity downgraded the evidence level and more well-designed studies with large samples are needed.Additional file 1:Search strategy for relevant studies. (DOCX 12 kb)


Type 2 diabetes mellitus (T2DM) is a global epidemic associated with increased health expenditure, and low quality of life. Many non-genetic risk factors have been suggested, but their overall epidemiological credibility has not been assessed.2 metric. We examined the presence of small-study effects and excess significance bias. We assessed the epidemiological credibility through a set of predefined criteria.We searched PubMed to capture all meta-analyses and Mendelian randomization studies for risk factors of T2DM. For each association, we estimated the summary effect size, its 95% confidence and prediction interval, and the IWe captured 86 eligible papers (142 associations) covering a wide range of biomarkers, medical conditions, and dietary, lifestyle, environmental and psychosocial factors. Adiposity, low hip circumference, serum biomarkers , an unhealthy dietary pattern , low level of education and conscientiousness, decreased physical activity, high sedentary time and duration of television watching, low alcohol drinking, smoking, air pollution, and some medical conditions  presented robust evidence for increased risk of T2DM.A healthy lifestyle pattern could lead to decreased risk for T2DM. Future randomized clinical trials should focus on identifying efficient strategies to modify harmful daily habits and predisposing dietary patterns. Type 2 diabetes mellitus (T2DM) ranks highly on the international health agenda as a global pandemic and as a threat to human health and global economies. The number of people with T2DM worldwide has more than doubled during the past 20 years . AccordiT2DM results from an interaction between genetic and environmental factors . Genes aSystematic reviews and meta-analyses of observational studies have indicated numerous risk factors for T2DM. However, the epidemiological credibility of these associations has not been appraised across the field. In the present work, we performed an umbrella review of the evidence across existing systematic reviews and meta-analyses of observational studies that examine any non-genetic risk factor for T2DM. We primarily aim to provide an overview of the range and validity of the reported associations of diverse environmental risk factors and biomarkers with T2DM. Furthermore, we assessed whether there is evidence for diverse biases and which of the previously studied associations have robust evidence.We conducted an umbrella review, i.e. a comprehensive and systematic collection and evaluation of systematic reviews and meta-analyses performed on a specific research topic using previously described and applied methodology –12.We systematically searched PubMed from inception until February 10, 2016 to identify systematic reviews and meta-analyses of observational studies examining associations of non-genetic risk factors with T2DM. We used the following search strategy: diabetes AND . Two independent investigators  retrieved and abstracted the full text of potentially eligible articles. We excluded meta-analyses that investigated the association between genetic polymorphisms and risk for T2DM; that included less than 3 component studies; that included studies with overlapping populations; that included studies using different units of comparison of the same exposure without transforming the effect estimates appropriately. We further excluded meta-analyses performing comparison between drug agents and subsequent risk for developing T2DM in population at high risk. When an association was covered by more than one meta-analyses, we kept the meta-analysis including the largest number of component studies and adequately presenting the study-specific effect estimates and sample sizes of component studies. We did not apply any language restrictions in our search strategy.Finally, in order to assess the causality of the associations between the reported risk factors and T2DM, we conducted an additional systematic search on PubMed to capture mendelian randomization (MR) studies for T2DM. This search algorithm used the keywords: “mendelian randomization” OR “mendelian randomisation”. MR studies were eligible if they studied T2DM and examined the potentially causal effect of a risk factor that was also included in our umbrella review. We excluded studies focused on impaired glucose tolerance, impaired fasting glucose or insulin resistance as outcomes.Two independent investigators  extracted the data, and in case of discrepancies consensus was reached. From each eligible article, we abstracted information on the first author, journal and year of publication, the examined risk factors and the number of studies considered. We also extracted the study-specific risk estimates  along with their 95% confidence interval (CI) and the number of cases and controls in each study. If a meta-analysis included multiple effect estimates from the same observational study using the same control group, we included only the effect estimate that corresponded to the largest sample size.From each eligible MR study, we extracted the first author and year of publication, the definition of outcome, the risk factor considered, the level of comparison for exposure, the genetic instrument used, the applied statistical approach, the sample size, the causal odds ratio and its 95% CI, the P-value for the association, and whether the authors claimed that a causal relationship exists. If an MR study used a genetic instrument based on a single variant and a genetic instrument based on polygenic risk score (PRS), we extracted the information from the PRS, as this approach is more powerful.For each meta-analysis, we estimated the summary effect size and its 95% CI using both fixed-effect and random-effects models. ,14 We al2 metric. I2 ranges between 0% and 100% and quantifies the variability in effect estimates that is due to heterogeneity rather than sampling error. [Between-study heterogeneity was quantified using the Ig error.  Values et distribution. [We assessed small-study effects using the Egger’s regression asymmetry test. ,19 A P <ibution. ,22 Excesibution.  In two mconvincing the associations that fulfilled all the following criteria: statistical significance per random-effects model at P <10−6; based on >1,000 cases; without large between-study heterogeneity (I2<50%); 95% PI excluding the null value; and no evidence of small-study effects and excess significance bias. Associations with >1,000 cases, P <10−6 and largest study presenting a statistically significant effect were graded as highly suggestive. The associations supported by >1,000 cases and a significant effect at P <10−3 were considered as suggestive. The remaining nominally significant associations (P <0.05) were considered as having weak evidence.We identified associations that had the strongest evidence and no signals of large heterogeneity or bias. We considered as For associations with convincing and highly suggestive evidence, we performed a sensitivity analysis limited to prospective cohort studies and nested case-control studies, and we examined whether there was a change in the level of epidemiological credibility. Also, we compared the findings from the meta-analyses of observational studies with the findings from MR studies.The statistical analysis and the power calculations were done with STATA version 12.0 and RStudio version 1.0.44.Our literature search yielded 7,303 papers, of which 86 papers met our inclusion criteria . FourteeThe 86 eligible papers examined 109 unique risk factors and 142 associations related to risk for developing T2DM. These associations covered a wide range of exposures: biomarkers (n = 25 associations), dietary factors (n = 53 associations), lifestyle factors and environmental exposures (n = 22 associations), medical history (n = 16 associations), metabolic factors and anthropometric traits (n = 15 associations), and psychosocial factors (n = 11 associations). The median number of cases per meta-analysis was 8,825 , and the median number of datasets was 10 . Only 7 meta-analyses included less than 1,000 T2DM cases.−6 (−6.One hundred and sixteen of 142 associations (82%) presented a statistically significant effect at P <0.05 under the random-effects model, whereas 46 associations had a statistically significant effect at P <10−6 . Fig 2 d2 >75%), and 50 associations (35%) had large heterogeneity estimates (I2 ≥50% and I2 ≤75%). The Egger’s test was statistically significant in 32 meta-analyses (23%), and 27 of them presented evidence for small-study effects. Thirty-nine meta-analyses (28%) had evidence for excess significance bias.Thirty-eight associations (27%) were very heterogeneous  for risk of T2DM. Low whole grains consumption, metabolically healthy obesity, increased sedentary time, low adherence to a healthy dietary pattern, high level of serum uric acid, low level of serum vitamin D, decreased conscientiousness, preterm birth, high consumption of sugar-sweetened beverages, high level of serum ALT, and exposure to high level of PM10 were associated with increased risk for T2DM and supported by convincing evidence.Eleven associations (8%) presented convincing evidence . Twenty-nine associations had suggestive evidence (20%), and 42 associations had weak evidence (30%) for risk of T2DM.Thirty-four associations (24%) were supported by highly suggestive evidence. The associations that were linked with a higher risk for T2DM and presented highly suggestive evidence were the following: high BMI , low educational status, gestational diabetes, increased processed meat consumption, high level of total and leisure-time physical activity, metabolically unhealthy obesity, psoriasis, low coffee consumption, high systolic blood pressure, high level of serum gamma-glutamyl transferase , metabolic syndrome, increased time of television watching, low hip circumference, late age at menarche, weight gain in early adulthood, weight gain after the age of 25 years, increased dietary heme iron intake, high level of serum C-reactive protein , low level of serum adiponectin (per 1 log μg/ml increase), low alcohol consumption, smoking , smoking cessation (new quitters vs. never smokers), major depressive disorder, bipolar disorder, high waist-height ratio, high waist circumference, high waist-to-hip ratio, and exposure to high level of NO−6 under random-effects model  were based on a combination of cross-sectional studies and cohort studies. In a sensitivity analysis limited to prospective cohort studies, the associations for sedentary time, hip circumference, and age at menarche remained highly suggestive . For psots model . In the We identified 22 MR studies assessing the causal effect of a risk factor that was included in our umbrella review . The medA causal effect was claimed for 4 risk factors graded as highly suggestive in our umbrella review: BMI, systolic blood pressure, serum gamma-glutamyl transferase, and waist circumference. A causal association was also claimed for birth weight, but a relatively small number of T2DM cases was included in this analysis. The observed effects for alcohol intake, coffee intake, serum CRP, serum ferritin, serum uric acid and serum vitamin D were not causal. Milk intake presented weak evidence in our analysis and an MR study did not show a causal effect. Serum adiponectin was graded as highly suggestive in our analysis, but the findings from MR studies were conflicting, and the largest MR study indicated absence of a causal effect.We performed a mapping of environmental factors and biomarkers examined for an association with T2DM in systematic reviews and meta-analyses. Overall, more than 100 associations were considered. We identified eleven associations supported by convincing evidence and thirty-four additional associations having highly suggestive evidence for risk of T2DM. These associations mainly pertained to comorbid medical conditions, lifestyle and dietary factors, as well as serum biomarkers.Even though more than one third of the associations examined various dietary factors, only six of them showed convincing or highly suggestive relationship with T2DM and the demonstrated effect sizes were modest. These factors were processed meat, whole grain products, healthy dietary pattern, sugar-sweetened beverages and dietary heme iron. Increased processed meat and sugar-sweetened beverages consumption are linked with other unhealthy lifestyle factors which showed highly significant association with T2DM, such as physical inactivity, increased BMI, smoking and unhealthy dietary patterns. ,24 The aModerate alcohol consumption has a protective effect against developing T2DM. This relationship could be explained by increased insulin sensitivity, lower fasting insulin resistance and lower glycated hemoglobin concentrations, which are induced by moderate amounts of alcohol. Moreover, moderate amount of alcohol drinking is a common feature of healthy diet pattern, who also lowered the risk for developing T2DM. Furthermore, coffee consumption lowers the risk for T2DM, which is attributed to the reduction of insulin resistance and the improvement of glucose metabolism. However, it is unclear whether this association is causal, given the findings of a recently published MR study .−54). Increased BMI, waist-height ratio, waist-hip ratio and waist circumference express the presence of increased intra-abdominal visceral fat, which disrupts insulin metabolism through release of serum free fatty acids. [Most of the associations yielded from our analyses were proxies of obesity and include body mass index (BMI), weight gain, and anthropometric characteristics . The observed association between BMI and T2DM demonstrated a large effect size and was highly significant , gamma-glutamyl transferase, C-reactive protein (CRP), uric acid, adiponectin, and vitamin D. High serum ALT and gamma-glutamyl transferase in patients with T2DM could be a manifestation of ongoing low-grade hepatic inflammation or hepatocellular damage, which is common in T2DM and metabolic syndrome. Among hepatic enzymes, ALT is the most specific indicator of hepatic pathology in non-alcoholic fatty liver disease and most closely related to liver fat accumulation. The presence of systemic inflammation is linked to β-cell dysfunction, leading to impaired glucose metabolism and the development of T2DM.  Both CRPFurthermore, our results indicated an inverse association between vitamin D level and risk for T2DM. It is unclear if this is a true association or the effect of adiposity as a potential confounder or intermediate factor. Obesity leads to storage of vitamin D in adipose tissue and to less sun exposure, on the grounds of limited mobility and accumulation of subcutaneous fat . All theThe association between smoking and T2DM has biological foundation because smoking is associated with central obesity and increased oxidative stress and inflammation, and eventually leads to insulin resistance and hyperglycaemia. However, residual confounding can be the case since smoking is often linked to other unhealthy lifestyle factors  and comorbidities. The increased risk of T2DM associated with smoking cessation in new quitters could be mediated by weight gain or be due to reverse causation because people who try to quit smoking are more likely to have preclinical conditions or high cumulative smoking exposure .Based on our assessment, adults delivered preterm presented a larger risk for development of T2DM during adulthood than adults delivered full-term. According to the “fetal origin of disease” hypothesis, the biological mechanisms that mediate this association could be explained through intrauterine growth restriction. Preterm newborns have low birth weight and they are prone to disrupted glucose metabolism in later life , which i10 and NO2, were found to have robust association with risk for T2DM. It has been suggested that air pollution causes elevated systemic inflammation and oxidative stress, whereas it increases the insulin resistance leading to abnormal glucose metabolism and elevated fasting glucose. [Two components of ambient air pollution, PMglucose. Furthermore, older age at menarche was associated with risk for T2DM. However, there are doubts whether it constitutes a genuine association. Observational studies found that this association is attenuated after adjustment for BMI in adulthood, suggesting that adult adiposity may mediate this association. The inverse association between age at menarche and BMI in adulthood could explain this finding. We presented an exposure-wide mapping of the meta-analyses on non-genetic risk factors for T2DM. Our umbrella review indicated that a very wide range of risk factors has been considered for T2DM. Compared to previously published umbrella reviews –10, therOur study has some caveats. First, the statistical test for small-study effects should be interpreted with caution in case of large between-study heterogeneity. Second, the observational studies did not often clearly report the sample sizes for the statistical analyses. Thus, the power calculations might be conservative, and the extent of excess significance bias is probably conservative. Furthermore, genetic instruments of the MR studies were not assessed, and power calculations for the MR studies could not be performed, because the percentage of variance explained often was not available. Consequently, the claims of MR studies should be interpreted with caution.Our paper identified several robust risk factors for T2DM. Our findings indicate specific strategies for public health interventions to reduce the future incidence of T2DM. Interventions for the promotion of physical activity and a healthy lifestyle and dietary pattern combined with interventions against the increased incidence of obesity could alleviate the projections for an increase of T2DM incidence in near future. However, these findings are based on observational data and should be interpreted with caution. Even though MR studies may support or not causality, the power of those studies could not be assessed. Therefore, randomized clinical trials and additional well-designed MR studies are needed to clarify which of these observations are causal associations. Also, these findings should be replicated by large-scale environment-wide association studies in various ethnic groups, and they could be used for the development of reliable risk prediction models in combination with known genetic polymorphisms.S1 File(DOC)Click here for additional data file.


RR = 0.75, 95% CI 0.64–0.89). In subgroup analysis by ovarian cancer subtypes, we found a statistically significant association between the dietary calcium  and the risk of epithelial ovarian cancer (EOC). This meta-analysis indicated that increased calcium intake might be inversely associated with the risk of ovarian cancer; this still needs to be confirmed by larger prospective cohort studies.Several epidemiological studies have evaluated the association between calcium intake and the risk of ovarian cancer. However, the results of these studies remain controversial. Thus, we performed a meta-analysis to explore the association between calcium intake and the risk of ovarian cancer. Pubmed, Embase and Web of Science were searched for eligible publications up to April 2017. Pooled relative risks (RRs) with 95% confidence intervals (CIs) were calculated using the random-effects model. Small-study effect was estimated using Egger’s test and the funnel plot. Among 15 epidemiological studies involving 493,415 participants and 7453 cases eligible for this meta-analysis, 13 studies were about dietary calcium intake, 4 studies about dairy calcium intake and 7 studies about dietary plus supplemental calcium intake. When comparing the highest with the lowest intake, the pooled RRs of ovarian cancer were 0.80 (95% CI 0.72–0.89) for dietary calcium, 0.80 (95% CI 0.66–0.98) for dairy calcium and 0.90 (95% CI 0.65–1.24) for dietary plus supplemental calcium, respectively. Dietary calcium was significantly associated with a reduced risk of ovarian cancer among cohort studies  and among case-control studies ( Ovarian cancers include three major histologic types: epithelial, sex cord/stromal and germ cell cancer. Approximately 90% of ovarian cancers are classified as epithelial ovarian cancer (EOC) ,2. ThereThe majority of cases are usually diagnosed at an advanced stage ,6, contrAs an important component in foods, calcium has been identified as being associated with many diseases, for instance, cardiovascular disease , stroke We followed the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) guidelines in this meta-analysis .A literature search was performed up to April 2017 for relevant available articles from the PubMed, Embase and Web of Science databases. We used the search terms “nutrition” OR “diet” OR “dietary” OR “calcium” in combination with (“ovarian” OR “ovary”) and (“neoplasm” OR “carcinoma” OR “cancer” OR “tumor”). We also reviewed the reference lists of the included studies for undetected relevant studies.The inclusion criteria were as follows: (1) a case-control or cohort study published as an original study; (2) the exposure of interest was calcium intake; (3) the outcome of interest was ovarian cancer; (4) relative risk (RR) with 95% confidence interval (CI) were available . (5) the most recent and complete study was selected if data from the same population had been published more than once.All studies were carefully searched and reviewed independently by two investigators. If the two investigators disagreed about the eligibility of an article, it was resolved by consensus with a third reviewer.The Newcastle-Ottawa quality assessment scale was used to assess the quality of the original studies. Quality of selection, comparability, and exposure or outcome of study participants are three major parameters. And a higher score represents better methodological quality.The following data were extracted from each study by two investigators independently: the first author, publication year, country in which the study was conducted, study design, follow-up duration, age range or mean age at baseline, sample size, number of cases, dietary assessment method, the most adjusted RR with 95% CI for the highest versus lowest category of the intake of calcium, and the covariates that were adjusted for in each study.I2 was adopted to assess the heterogeneity between studies  [Pooled measure was calculated as the inverse variance-weighted mean of the logarithm of RR with 95% CI to assess the strength of association between calcium intake and the risk of ovarian cancer. The random effect model was used to combine study-specific RRs (95% CIs) . The I2 ctively) . Meta-rectively) . In the ctively) .p-values) were two-sided with p ≤ 0.05 considered statistically significant.All statistical analyses were performed with STATA version 12.0 . All reported probabilities  and Europe . When stratified by study design subtype, a statistically significant effect of dietary calcium on ovarian cancer risk was observed both among case-control studies  and cohort studies . For subgroup analysis stratified by dietary assessment method, the inverse association was also statistically significant in validated FFQs group  and in no-validated FFQs group . The remaining results of subgroup analyses are shown in In subgroup analysis stratified by continent in which the studies were conducted, dietary calcium was significantly associated with decreased ovarian cancer risk among studies conducted in North America  involving 317,995 participants and 3780 cases were included. Two studies revealed a significant association between dietary plus supplemental calcium intake and the risk of ovarian cancer, while the other five studies found no association. The pooled RR of ovarian cancer was 0.90  involving 155,859 participants and 2330 cases were included, and the pooled RR of ovarian cancer was 0.80 , continent (p = 0.296), study design (p = 0.394), whether adjusted for energy intake (p = 0.319), parity (p = 0.584), oral contraceptive use (p = 0.896), tubal ligation (p = 0.196) and dietary assessment method for calcium intake (p = 0.033). The results showed that the dietary assessment method contributed to the between-study heterogeneity.Univariate meta-regression with covariates was conducted to explore the source of heterogeneity. In analysis of dietary calcium with risk of ovarian cancer, we performed univariate meta-regression with the covariates of sample size (p = 0.150), study design (p = 0.029), dietary assessment method (p = 0.190) and whether adjusted for parity (p = 0.874), oral contraceptive use (p = 0.190) and tubal ligation (p = 0.401). The results showed that the study design contributed to the between-study heterogeneity.In analysis of dietary plus supplemental calcium with risk of ovarian cancer, we performed univariate meta-regression with the covariates of sample size , dietary plus supplemental calcium (p = 0.501) and the risk of ovarian cancer. The funnel plot of the analysis of dietary calcium and the risk of ovarian cancer is shown in the Egger’s test showed no evidence of a significant small-study effect for the analyses between the consumption of dietary calcium  ,22. The Pheterogeneity = 0.124) for cohort studies and 0% (Pheterogeneity = 0.508) for case-control studies, respectively. Additionally, low between-study heterogeneity was found in the analysis of dietary calcium and risk of ovarian cancer. In meta-regression, we found that the dietary assessment method was the contributor to between-study heterogeneity. In subgroup analysis by studies adjusted for dietary assessment method or not, the heterogeneity decreased to 20.5% (Pheterogeneity = 0.254) and 0% (Pheterogeneity = 0.875), respectively. The results were also consistent with the overall pooled RR, indicating that our results were stable and reliable.Between-study heterogeneity is common in meta-analysis , and it There are several strengths in present meta-analysis. First of all, this study included a large number of participants and cases, allowing a much great possibility of reaching reasonable conclusions. Second, most of the studies had adjusted for potential confounders, such as age, energy and oral contraceptives, strengthening the credibility of the results. Third, most of the included studies used validated FFQs, ensuring the accuracy of dietary assessment. Fourth, a statistically significant inverse association between dietary calcium and ovarian cancer risk was found in cohort studies, indicating a potential causal relationship between them. Fifth, in cumulative meta-analysis by publication year, the significant result persisted and the CI became increasingly narrower.However, several limitations of the study should also be considered. First, potential confounders adjusted for in each study were different and it might affect the results to some extent. For example, parity and tubal ligation were adjusted for in some studies, while they not adjusted for in other studies. In addition, residual confounding should be of concern. Second, most studies just used a complete dietary assessment, and it could not reflect change in diet for a long time. Third, the standards of lowest calcium intake were inconsistent among studies, which might influence the result. Fourth, case-control studies, which are prone to recall and select biases, were included in our meta-analysis, and the results would be partly affected. Fifth, the number of included studies for dairy calcium was too small (only four studies). Although the result of dairy calcium with risk of ovarian cancer was statistically significant, it was different from the result reported by Genkinger et al. in 2006 . Sixth, In summary, this meta-analysis revealed that dietary calcium and dairy calcium, but not dietary plus supplemental calcium, may reduce the risk of ovarian cancer. Increasing intake of dietary calcium should be advocated for the primary prevention of ovarian cancer. Better designed prospective cohort studies are needed to explore the association between calcium intake and ovarian cancer risk.


In recent years, accumulating evidence has supported the hypothesis that lower vitamin D status is associated with several known risk factors of stroke. However, the relationship between vitamin D and stroke is still uncertain. To explore if there was an association between vitamin D status and the risk of stroke, a systematic review and a meta-analysis were conducted by searching three databases: Pubmed, Embase, and the Cochrane Library. Following the application of inclusion and exclusion criteria, the relative risk estimates of all the included studies were pooled together to compare the risk of stroke between the lowest and the highest category of vitamin D. The Newcastle–Ottawa Scale (NOS) and the Cochrane Risk of Bias Tool were used to assess the risk of bias, and the publication bias was detected by using a funnel plot and Egger’s test. Nineteen studies were included and the pooled relative risk was 1.62 (95% CI: 1.34–1.96). Further analysis found that vitamin D status was associated with ischemic stroke , but not with hemorrhagic stroke . In conclusion, our meta-analysis supported the hypothesis that lower vitamin D status was associated with an increased risk of ischemic stroke. Further studies are required to confirm this association and to explore the association among different subtypes. Stroke is the second leading cause of death among people aged 60 years and above , and is Vitamin D, which is classically known as a protective factor in bone metabolism, in recent years has been reported to play a vital role in cardiovascular health ,5,6. An In addition, a recent study that combined the data of three population-based studies reported no significant association between vitamin D status and the incidence of hypertension and stroke ,13. MoreThis systematic review and meta-analysis was based on a protocol published in the 2017 PROSPERO register (CRD: 42017076416), and the report followed the PRISMA  guidelines.A literature search was performed using three databases: Pubmed, Embase, and the Cochrane Library, from inception through to 30 September 2017. When searching, Medical Subject Headings (MeSH) terms and free text words were combined. Detailed search strategies for each database are shown in Published in English.Conducted among humans.Being a case control study, a cohort study, or a randomized controlled trial (RCT).Studying the association between vitamin D status  and the risk of incident stroke.Containing information about contingency tables, the odds ratio (OR), or the risk ratio (RR), as well as their 95% confidence interval (CI).The inclusion criteria were as follows: Repeating publications or duplicates.Insufficient data for analysis.The exclusion criteria were as follows:Two reviewers independently completed the study selection and data extraction. Disagreements in the study selection and data extraction process were resolved by discussion. The study selection was conducted according to a strict adherence to the inclusion and exclusion criteria. If several studies used the same data source for analysis, the one with the largest sample size was included and the others were excluded. As for the data extraction, information including the first author, publication year, journal, country or district, study design, sample size, OR or RR and their 95% CI, adjusted variables, different subtypes of stroke, and the vitamin D status were extracted. If several models were reported, the estimates in the most adjusted models were extracted.Two different assessment tools were used for different study designs. The Newcastle–Ottawa Scale (NOS) was used to assess the quality of non-randomized studies, including case control studies and cohort studies. The scale includes eight items with one or two points for each item. Studies with zero to three points were considered with a high risk of bias. As for the RCTs, the Cochrane Risk of Bias Tool with eight items was used to assess the risk of bias, with one point for each item. Studies with zero to two points were considered with a high risk of bias.Q test and I2 statistics were used to assess the heterogeneity among the included studies. If no significant heterogeneity existed (I2 statistics less than 50%), the pooled estimate and the 95% CI were calculated using a fixed-effect model. If significant heterogeneity was present (I2 statistics no less than 50%), then a random-effect model was adopted. Review Manager 5.3  was used for data analyses.Risk ratios between the highest and the lowest category of vitamin D in prospective studies and the odds ratios in retrospective studies were used to perform the data synthesis. The pooled relative risk and the 95% CI of the included studies were calculated using the inverse variance method, with ORs of retrospective studies being approximately considered as RRs. Two kinds of sensitivity analyses were performed. Studies with extraordinary estimates of the association (vitamin D and stroke), and studies with scores of less than 7 in the assessment of risk of bias were respectively removed. After the removal, the pooled relative risk was calculated again using the remaining studies. If the pooled estimates changed from significant to insignificant (or vice versa), or the pooled estimates changed from more than 1 to no more than 1 (or vice versa), we then considered that influential outliers existed among the included studies.Further subgroup analyses were also performed according to study designs , sample size (more or no more than 1000), subtypes of stroke (ischemic stroke or hemorrhagic stroke), and vitamin D indicators (circulating vitamin D or vitamin D intake), to explore if the association differs in different subgroups and to locate the origin of the heterogeneity.A funnel plot was used to visually assess the publication bias. Egger’s test (significant level = 0.05) was conducted to quantitatively explore the possible publication bias, using STATA 14.0 .The meta-analysis and report of the results were based on the PRISMA checklist and the details are shown in A total of 2319 studies were retrieved from the three databases, of which 19 studies were finally included for analysis. p value of the Q test was less than 0.01 and the I2 statistic = 79%, indicating that heterogeneity existed among the included studies.We calculated the relative risk between the lowest with the highest categories of vitamin D in all of the included studies. The pooled relative risk was 1.60  also suggested that there might be small-study effects in our study.The dissymmetry of the funnel plot shown in  implied Our meta-analysis included 19 studies that explored the association of vitamin D status with the risk of stroke. Comparing the lowest and highest category of vitamin D using the random-effect model, the pooled relative risk for all included studies was 1.62 [95% of confidence interval (CI): 1.34–1.96], which indicated an inverse association between vitamin D status and the risk of stroke. Further subgroup analyses were performed and found that study designs and sample size did not affect the association between vitamin D status and the risk of stroke. The subgroup analyses indicated that vitamin D status was associated with ischemic stroke, but not with hemorrhagic stroke. In addition, we did not find a significant association between vitamin D intake and stroke. Our results were generally consistent with two earlier meta-analyses. Wang et al.  reportedMany studies have provided evidence for possible mechanisms to explain the effect of vitamin D on stroke. Although vitamin D is known for its regulation of bone health, vitamin D receptor (VDR) was expressed in most human tissues and cells . More imIn the subgroup analyses, the estimated effect size of the case control studies showed a stronger association between vitamin D status and stroke than that of the prospective studies, which might be mainly due to the two case control studies with distinct and higher estimated ORs ,29. SincOur study did not support the hypothesis that vitamin D status had an effect on hemorrhagic stroke. As mentioned above, vitamin D can inhibit the development of thrombosis, which may provide a rational explanation for the relationship between vitamin D and ischemic stroke. However, this explanation was not suitable for hemorrhagic stroke as hemorrhagic stroke was not based on the development of thrombosis. Nevertheless, vitamin D might induce hemorrhagic stroke through other mechanisms, such as inflammation and the RAS. Since the number of original studies exploring the association between vitamin D status and stroke was quite limited, more studies are needed to further investigate the association between vitamin D and the risk of hemorrhagic stroke. The estimated relative risk in the vitamin D intake group was insignificant. Firstly, only three studies that reported the vitamin D intake were included in our analysis, resulting in the limited power to identify the association. In addition, the measurement of vitamin D intake might differ among different studies. Further studies on the intake of vitamin D will provide better evidence for estimating.Recently, a Mendelian randomized study was conducted to explore the causal relationship between serum vitamin D and the risk of ischemic stroke . AlthougIn the current study, we updated older meta-analyses using recently published studies. In addition, to explore the heterogeneity among these studies, further subgroup analyses were conducted. This study found that vitamin D might play different roles  in different subtypes. However, our study has several limitations. Firstly, we estimated the association between vitamin D status and stroke by comparing the lowest and highest category of vitamin D. However, the category standards were not the same among our included studies. This heterogeneity among studies might introduce unknown influence on the pooled estimates. To resolve this problem, when the heterogeneity was high, the random-effect model was used to calculate the pooled estimate. Second, the funnel plot and Egger’s test indicated that publication bias might exist in our studies. Since this study used published studies as data sources, it was difficult to prevent publication bias. Thirdly, as the original studies included different confounder sets in the fully adjusted models, heterogeneity could be introduced into the meta-analysis. However, a previous study reported that choosing effect estimates based on the fully adjusted models would not substantially change the results of the meta-analysis, compared to when using models that control only the most common confounders (such as sex and age) . TherefoLower vitamin D status was associated with an increased risk of stroke by synthesizing the data of nineteen relevant studies. This association remained in the ischemic stroke group, while the relationship between vitamin D status and hemorrhagic stroke was insignificant. Higher circulating vitamin D was a significant protective factor for stroke, while the association between vitamin D intake and the risk of stroke remained to be further explored. More studies, especially randomized controlled trials, are needed to verify this association and explore the prevention effectiveness of vitamin D supplement.


CYP2E1 polymorphisms have been reported to influence individual’s breast cancer susceptibility as a phase I enzyme, but the results of these previous studies remain controversial. We performed a comprehensive meta-analysis to assess their association.A comprehensive search of literature included in various databases , published before August 2016, was performed. Odds ratios (ORs) with 95% confidence intervals (CIs) calculated in fixed or random-effects models were used to estimate the strength of the associations between three polymorphisms of CYP2E1 and breast cancer susceptibility. Subgroup analysis, sensitivity analysis and test for publication bias were also performed. A total of 11 separate comparisons involving 4311 cases and 4407 controls were included in the meta-analysis.C>T, CYP2E1*5 Rsa I/Rst I (c1/c2) and BC risk. For CYP2E1*6 Dra I (D/C) polymorphism, a significantly increased BC risk in the overall population was found in genetic model D/C vs. D/D  and C/C + D/C vs. D/D , together with subjects who have at least one C allele . Similar results were also found in subgroup analyses in Caucasians of these three comparison models.Our result showed that there was no significant association between the two common polymorphisms CYP2E1 rs2031920 Dra I (D/C) variation significantly associated with the risk of BC. Individuals with D/C and C/C + D/C genotypes or carried at least one C allele of CYP2E1*6 Dra I (D/C) polymorphism had a significant higher susceptibility to develop BC.The present meta-analysis suggests that CYP2E1*6 The online version of this article (doi:10.1186/s12935-016-0371-9) contains supplementary material, which is available to authorized users. Worldwide, breast cancer (BC) represents the leading cause of female gynecological cancer death. Its estimated deaths 189,000) were reported almost equal to the estimated number of deaths from lung cancer  89,000 we, but theN-nitrosamines in cigarettes [S-transferases, N-acetyltransferases, and sulfotransferases, interacting with carcinogen exposure, may modified one’s susceptibility to cancer [N-nitrosamines, altering its activity has been suggested might link to the development of BC [Cytochrome P4502E1 (CYP2El), a member of Cytochrome P450 (CYP) super-family which involved in the metabolism of many endogenous and exogenous substance, is of pivotal importance in metabolizing ethanol and low-molecular-weight carcinogens such as garettes . Alcoholgarettes . The relgarettes . Furthero cancer . Among thttp://www.ncbi.nlm.nih.gov/SNP). However, only several common mutations were extensively investigated as they might alter the activity of CYP2E1 [G>C and rs2031920 C>T were two key SNPs among them, with the former one associated with Pst I restriction enzyme site and the later one with Rsa I restriction enzyme site, their linkage disequilibrium also lead to the CYP2E1*5 haplotype and form three types of distinct genotypes: (1) c1/c1(Rsa I+ Pst I−), homozygous of normal alleles; (2) c1/c2, heterozygous; (3) c2/c2 (Rsa I− Pst I+), homozygous alleles after nucleotide exchanged [C/C, C/D and D/D [CYP2E1 is located at chromosome 10. So far, more than 100 single nucleotide polymorphisms (SNPs) have been found . While the most recently study by Chong et al. [The relationship between above CYP2E1 polymorphisms and BC has been investigated by various studies, however, presenting conflict results. One studies conducted on patients suffering from primary unilateral BC demonstrated the absence of any association between CYP2E1*5 polymorphisms with BC, no matter in premenopausal or postmenopausal women . Howeverg et al. . indicatA comprehensive search of literature listed in various databases , published before August 2016, was performed using the following key words ‘breast cancer’ or’ ‘breast carcinoma’, ‘polymorphism’ or ‘variant’ and ‘mutation’, all combined with Medical Subject Heading (MeSH) term ‘CYP2E1’. The eligible studies were retrieved, and their reference lists were screen by hand to find every relevant paper. No any restriction such as time and language was made during the searches, as well as attempts to obtain unpublished studies.In this study, we performed the meta-analysis according to the proposal of Meta-analysis of Observational Studies in Epidemiology group (MOOSE) . The eliData extraction from each eligible study was conducted by two independent investigators, which included: (1) the first author’s name; (2) year of publication; (3) study region or country; (4) ethnicity; (5) cancer confirmation; (6) sample size (both cases and controls); (7) source of control (together with matching criteria); (8) polymorphisms of CYP2E1; (9) genotyping method; (10) genotype distribution in cases and controls and whether P value of the control population consistent with the Hardy–Weinberg equilibrium (HWE). In the event of different results, discussion was conducted to solve the discrepancies. When a study reported the results on different CYP2E1 polymorphisms, we treated them as separate studies in our meta-analysis.To evaluate the quality of the included studies, a set of predefined criteria originally proposed by Thakkinstian et al. . was use2 statistics. According to the presence (PQ < 0.1 or I2 ≥ 50%) or absence (PQ ≥ 0.1 and I2 < 50%) of heterogeneity, different models would be used to calculate the pooled ORs, with the former situation using DerSimonian–Laird random-effects model while the later using Mantel–Haenszel fixed-effects model. If heterogeneity existed, Galbraith plot analyses would be carried out to investigate the sources of heterogeneity among studies. Then, subgroup analysis by ethnicity would be performed to address possible effects of these polymorphisms on different population. To assess the stability of the results, sensitivity analysis was performed by sequential omission of individual studies, especially studies whose genotype frequencies in the control populations were deviated from HWE, as they may generate bias. HWE in the control group population, if not reported in the original article, would be tested via a goodness-of-fit Chi square test. Finally, for each polymorphism, the Begg’s funnel plots and Egger’s linear regression test was used to test the publication bias (P < 0.05 indicated a significant publication bias). All analyses were performed with Stata software  and all P values were two-sided.The association of each CYP2E1 polymorphisms with breast cancer risk was estimated by calculating pooled ORs and 95% CIs under different comparison models, including additive models, recessive model, and dominant model. Firstly, the heterogeneity between studies would be assessed by the Q test and IC>T, CYP2E1*5 Rsa I/Pst I and CYP2E1*6 Dra I) and breast cancer risk were included in our meta-analysis [There were 47 published articles relevant to the search terms. By browsing the title and abstract, 25 studies were excluded because of obvious irrelevance. After a careful full-text review of the remaining 22 studies, a further 13 articles were removed: 7 were reviews; 2 were not case–control studies; 2 focus on other SNPs of CYP2E1 or other cancer; and the rest 2 did not report sufficient data. Additional eligible studies were not found through manual search of the reference lists. Consequently, night case–control studies focus on three CYP2E1 polymorphisms  [Rsa I/Pst I polymorphism (three in Asian and one in Arab) [Dra I polymorphism  [C>T polymorphism while others were all reported or calculated consistence with HWE. All studies included met quality criteria ranging from 9 to 14, hence two studies were regarded as low-quality and night was high-quality. Basic characteristics of all eligible studies were listed in Table Of all the selected articles, three studies consist of 1915 cases and 1793 controls evaluated the association of CYP2E1 polymorphisms and breast cancer risk in rs2031920 ulation) , 22, 23;in Arab) , 25; then Asian) . Among tC>T polymorphism was not associated with BC risk in all genetic models in the overall populations: (1) T vs. C ; (2) T/T vs. C/C ; (3) C/T vs. C/C ; (4) T/T + C/T vs. C/C ; (5) T/T vs. C/T + C/C   polymorphism, we still failed to identify any significant association with BC susceptibility: (1) c2 vs. c1 ; (2) c2/c2 vs. c1/c1 ; (3) c1/c2 vs. c1/c1 ; (4) c2/c2 + c1/c2 vs. c1/c1 ; (5) c2/c2 vs. c1/c2 + c1/c1   variation, our result indicated a significant increased BC risk in genetic model D/C vs. D/D  and C/C + D/C vs. D/D  . When stratified by ethnicity, similar results were also found in Caucasians in these three comparison models. Details were present in Table With regard to the CYP2E1*6 T/T vs. C/C and T/T vs. C/T + C/C comparison models of rs2031920 C>T polymorphism, Galbraith plot analyses were carried out to detect the possible source of heterogeneity. However, as show in Fig. T/T vs. C/C model nor T/T vs. C/T + C/C model, indicating that the studies included in both two comparison models were not contributors to the heterogeneity.As significant heterogeneity was found in the C>T variation, sensitivity analysis were performed to see if any single study would greatly influenced the estimates of overall risk, our result showed that the pooled ORs did not materially altered with or without these studies (data not shown). But we could not conduct sensitivity analysis for T/T vs. C/C model and T/T vs. C/T + C/C model due to a limited study number in these two models (only three studies were included and one of them did not provided data in T/T genotype).Considering that there were two studies whose genotypes inconsistent with HWE in rs2031920 To assess possible publication bias, Begg’s funnel plots and Egger’s tests were performed simultaneously. The funnel plots were symmetrical in all genetic models of three CYP2E1 polymorphisms, indicating no significant publication bias existed in all the articles included. Egger’s test, with all the P value larger than 0.05, also revealed no evidence of publication bias in our meta-analysis Fig. .Fig. 4BeN-nitrosaminesan and alcohol, was in different activity among individuals [Breast cancer, of which heredity explains approximately 10–15% of the cases, with only 5% can be clarified by known genetic polymorphisms such as BRCA1 and BRCA2 . Such faividuals . It has C>T polymorphism and BC. The haplotype CYP2E1*5, which consist of SNP rs2031920 C>T and 3813867 G>C, also failed to identify any significant association with BC risk. Single SNP rs3813867 G>C was also taken into consideration, but further analysis was not carried out due to the suboptimal study numbers (n = 1) [C>T SNP and BC risk were consistent with those studies included in the present meta-analysis, all indicating an insignificant relationship between them; and the overall results of CYP2E1*5 polymorphism and BC were also accordance with half of those included, though one of the rest observed a decreased risk of BC while another revealed an increased risk. Taken together, it may be concluded that CYP2E1*5 polymorphisms are not associated with BC risk in the overall population.However, in the present study, no significant association was found between SNP rs2031920  (n = 1) . Such in (n = 1) . NeverthD/C and C/C + D/C genotype had a significantly higher risk of BC compared to those with the D/D genotype, similar increased result was also found in the C allele carriers when compared with the D allele carriers, especially in Caucasian population. These results suggested that polymorphism in CYP2E1*6 could be a risk factor for BC development. But such result was inconsistent with those of the original studies, of which all suggested no significant relationship between any comparison model of CYP2E1*6 variation and BC development.Interestingly, when considering the CYP2E1*6 polymorphism, our study found that individuals with the c1/c2 genotype frequencies between BC cases and controls. However, after taking the drinking situation into consideration, a 1.9-fold increasing risk for developing BC was found when comparing the ‘ever’-drinking women with the c2 mutation to the non-drinkers with the c1/c1 mutation. Another study, investigated lifetime passive cigarette smoke exposures together with genetic variants and BC risk in women who had never smoked, found that interaction between passive smoke exposure and CYP2E1*6 AA/AT (namely CC/CD) polymorphism could significant increased breast cancer risk among premenopausal women [Actually, our results should explain with caution as there is increasing evidence that metabolizing enzymes do not act alone. In the study carried out by Choi et al. . that exal women . In sum,To our knowledge, this is the first meta-analysis carried out to date to evaluate the role of CYP2E1 polymorphisms in breast cancer susceptibility. Despite the findings mentioned above, this study had several limitations. First, we haven’t taken the gene-environment interaction into consideration. As is known to all, apart from genetic factors, smoking status and alcohol consumption are important risk factors for BC; however, we could not conduct subgroup analyses stratified by environmental exposure due to the limited information on our included studies. Second, the overall results of our study were based on crude ORs, but a more precise evaluation should be adjusted for the know risk factors such as age and menopause status. Third, the number of studies included in this study is relatively small, with three or four studies for each polymorphism, which may lead to low statistical power and prevent us from exploring a real association of the CYP2E1 polymorphism and BC risk. Fourth, because no attempts were made to access unpublished studies and studies in languages other than English, publication bias may exist, though results of our Begg’s funnel plot and Egger’s test did not reveal any publication bias. Fifth, as most studies were conducted in Asian and Caucasian population, the relative lack of ethnic diversity demands for further studies.Dra I (D/C) polymorphism might be associated with increased BC risk, individuals with D/C and C/C + D/C genotypes or carried at least one C allele of CYP2E1*6 Dra I (D/C) polymorphism had a significant higher susceptibility to develop BC, in Caucasians, particularly. Whereas, no any significant relationship between CYP2E1*5 Rsa I/Rst I (c1/c2), rs2031920 C>T polymorphisms and BC risk was found.Aside from the above limitations, this meta-analysis suggests that CYP2E1*6


Epidemiological studies suggest that antidepressants use may increase the risk of cataract, but the results are inconclusive. We aimed to examine this association by performing a systematic review and meta-analysis.Relevant studies were identified by searching PubMed and Web of Science databases through June 2017. We included studies that reported risk estimates for the association between antidepressants use and cataract risk. A random-effects model was used to calculate the summary odds ratio (OR) with its 95% confidence interval (CI).P < 0.001, I2 = 92.2% for SSRIs, P = 0.026, I2 = 67.5% for SNRIs, and P = 0.092, I2 = 58.0% for TCAs).We identified seven studies of antidepressants use and risk of cataract involving 447,672 cases and 1,510,391 controls. Overall, the combined ORs (95% CIs) of cataract for selective serotonin reuptake inhibitors (SSRIs), serotonin noradrenalin reuptake inhibitors (SNRIs), and tricyclic antidepressants (TCAs) were 1.12 (1.06–1.19), 1.13 (1.04–1.24), and 1.19 (1.11–1.28), respectively. A certain degree of heterogeneity was observed across studies (This meta-analysis provides evidence of a significant positive association between antidepressants use and risk of cataract. Because of the heterogeneity and limited eligible studies, further prospective studies are warranted to confirm the preliminary findings of our study. Cataract is defined as partial or complete loss of transparency of the crystalline lens and is considered the primary cause of vision loss worldwide . The higRecently, emerging epidemiological studies have focused on the risk of cataract formation of antidepressants. Two population-based studies from Canada  and the Given the inconsistency and conflict of the existing literature and the insufficient statistical power of individual studies, we performed the present meta-analysis based on all eligible epidemiological studies that provided data on the association of antidepressants use with cataract risk.We performed this meta-analysis in accordance with the Meta-Analysis of Observational Studies in Epidemiology guidelines . A systeStudies included in this meta-analysis met the following criteria: 1) had cohort, nested case-control or case-control study design; 2) the exposure of interest was antidepressants use, including SSRIs, TCAs, serotonin noradrenalin reuptake inhibitors (SNRIs), and so on; 3) the endpoint of interest was cataract incidence; and 4) the risk estimate and the corresponding 95% confidence interval (CI) were reported. If multiple studies used the same population, we included the study with the largest sample size.We extracted the following data using a standardized data-collection form: last name of the first author, publication year, study region, number of cases and controls, method of exposure and endpoint assessment, types of antidepressants, risk estimates from the most fully adjusted model and the corresponding 95% CIs, matched or adjusted potential confounders.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). Two authors (QD and YF) independently performed the literature search, study selection, data extraction, and quality assessment. Any disagreements were resolved by consensus.We assessed the quality of individual studies using Newcastle-Ottawa Scale (NOS), a 9-star system which consists of three dimensions: selection (four items), comparability (one item), and exposure/outcome (three items) .The multivariable-adjusted odds ratios (ORs) for each study and the pooled ORs for the any exposure versus none of antidepressants are presented in Fig. The results of subgroup analysis according to individual antidepressant drugs are presented in Fig. As only seven studies were included in the present meta-analysis, Begg’s test  and EggeThis meta-analysis of seven eligible studies involving 447,672 cases and 1,510,391 controls supports a significant positive association of SSRIs, SNRIs and TCAs use with risk of cataract. To the best of our knowledge, this is the first systematic review and meta-analysis aimed to evaluate the relationship between antidepressants use and risk of cataract development.Heterogeneity is often a concern in a meta-analysis. In the present study, obvious heterogeneity was observed among most analyses, which was partially explained by the following factors: study design was different. Although most included studies were performed in Western countries, population characteristics still varied in genetic and environmental background, antidepressants use, and matched or adjusted confounders.Several mechanisms may be involved in the positive association of antidepressants use with cataract risk. In animal models, serotonin has been reported to play a crucial role in lens transparency . ElevateOur study had some important strengths. Considering individual studies had limited statistical power, this meta-analysis of seven studies involving a large number of cases and controls improved the power to detect a potential association and provided more robust estimates. Most of the original studies matched or adjusted a series of variables, which greatly reduced the likelihood of confounding bias.Potential limitations of our study should be considered. First, the number of eligible studies was limited, especially in some subgroup analyses, which might influence the reliability of the results. Second, significant heterogeneity was observed among included studies, which might distort the conclusion of our study. Third, a certain degree of publication bias was observed. Gray literature  was difficult to find and studies with null results were less likely to be published. Finally, random misclassification of antidepressants might influence the results.Use of antidepressants, including SSRIs, SNRIs and TCAs, is associated with an increased risk of cataract development. Considering the huge heterogeneity and limited included studies, further large well-designed prospective studies are warranted to confirm the preliminary findings of our study.


P < 0.001). In addition, subgroup analyses performed in Western population also confirmed the significant inverse relationship between PD and risk of lung cancer . In the subgroup analysis, a reduced risk of lung cancer in PD patients from Western population was consistent regardless of study design, gender, or study quality. In conclusion, PD patients were significantly associated with a reduced risk of lung cancer in Western population. The relationship between them in Asian population needs to be confirmed by future studies.Recently, growing evidence has revealed the significant association between Parkinson's disease (PD) and cancer. However, controversy still exists concerning the association between PD and lung cancer. A comprehensive article search for relevant studies published was performed using the following online databases: PubMed, Web of Science and Embase up to August 31, 2016. The pooled risk ratio (RR) and their 95 % confidence intervals (CI) were calculated using the method of inverse variance with the random-effects model. Fifteen studies comprising 348,780 PD patients were included in this study. The pooled result indicated that patients with PD were significantly associated with a decreased risk of lung cancer (RR: 0.53, 95% CI: 0.41−0.70,  Parkinson's disease (PD) is the second most common neurodegenerative disorders characterized by the selective dopaminergic cell loss in the substantia nigra, and the prevalence of PD of the entire population is about 0.3% in industrialized countries . Most PDIt is all to know that cancer is a disease characterized by cell infinite proliferation and lack of apoptosis. Recently, growing evidence has revealed the significant association between PD and cancer , 3, and In this study, we included the cohort and case-control studies to perform a meta-analysis which aimed to explore the association between PD and risk of lung cancer.A total of 371 potentially relevant studies were identified in the database search. After exclusion of 296 duplicated studies, there were remain 130 studies. After title and abstract review, 97 studies were excluded. After full article review for the remaining 33 studies, 7 duplicated studies which were performed in the same population and 11 studies which did not meet inclusion criteria were excluded . FinallyThese studies were published between 2000 and 2016, and contained 348,780 patients with PD. Among these 15 studies, 2 studies were performed in Asian population (1 in Japan and 1 in Taiwan), while 13 studies were conducted in Western population . The baseline characteristics of each study are shown in Table P < 0.001, I2 = 94%, Figure The result of 12 cohort studies and 3 case-control studies indicated that the pooled RR of lung cancer in PD patients versus control patients was 0.53 , but it could not be found in before PD group  cohort and/or case-control study evaluating the relationship between PD and risk of lung cancer; (2) an estimate of association  with measures of variation  was provided; (3) published study in English. When duplicated studies were based on the same population, only the most informative study was included , 18. CasData of each study was extracted independently by two authors (Xin Xie and Xiaoguang Luo) according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines  33], an, an33], Q test and I2 statistics [The pooled RR and their 95% confidence intervals (CI) were calculated using the method of inverse variance with the random-effects model. Statistical heterogeneity was evaluated using Cochran's atistics . Begg's atistics , 37. Cumatistics .P value was less than 0.05.STATA software  was used to perform the data analyses in this study. It was considered statistically significant when


CYP2E1) gene on the risk of squamous cell carcinoma (SCC), a meta-analysis was performed.To explore the genetic effect of rs2031920 and rs3813867 polymorphisms within the cytochrome P450 2E1  = 0.67], carrier T vs. carrier C , TT vs. CC , CT vs. CC , CT + TT vs. CC . Similarly, a decreased SCC risk was observed for the rs3813867 G/C polymorphism in the allele, carrier, homozygote, dominant, and recessive models of overall SCC meta-analysis and “ESCC” subgroup analysis  and in all genetic models of “Asian” and “population-based control (PB)” subgroup analysis . Additionally, for the rs2031920/rs3813867 haplotype, a decreased SCC risk was also detected in the overall SCC meta-analysis under the allele, carrier, homozygote and dominant model  and the subgroup analysis of “PB” under the allele, carrier, and dominant models .After the process of database searching and screening, a total of 32 case–control studies with 7435 cases and 10,466 controls were ultimately included in our meta-analysis. With regard to the rs2031920 C/T polymorphism, in comparison to controls, a reduced risk in cases of esophageal squamous cell carcinoma (ESCC) was detected for the models of allele T vs. allele C [YP2E1 rs2031920 C/T polymorphism and “C” allele carrier of the rs3813867 G/C polymorphism as protective factors for ESCC patients, especially in Asian populations.Our meta-analysis supports the “T” allele carrier of the C CYP2E1) gene in Homo sapiens is located on chromosome 10 and is responsible for encoding a membrane-bound CYP2E1 protein, an important member of the human cytochrome P450 system  were obtained, as presented in Fig. 61, CochrP value of association test > 0.05).A meta-analysis of rs2031920 and SCC risk was conducted on the allele model , carrier model (carrier T vs. carrier C), homozygote model (TT vs. CC), heterozygote model (CT vs. CC), dominant model (CT + TT vs. CC), and recessive model (TT vs. CC + CT). As shown in Table P = 0.025, OR = 0.67), carrier T vs. carrier C , TT vs. CC , CT vs. CC , CT + TT vs. CC , but not TT vs. CC + CT (P = 0.770). Figure CYP2E1 gene seems to be linked to ESCC risk.Moreover, we conducted a statistical analysis of the subgroup of race (Asian/Caucasian), SCC type (HNSCC/ESCC/LSCC), control source (PB/HB), and HWE (Y/N). As shown in Table CYP2E1 rs3813867 and SCC risk were detected in the overall SCC meta-analysis and subgroup analysis of “ESCC” and “Y”  under all genetic models , only apart from the heterozygote model (P = 0.150). A decreased SCC risk was also detected in the subgroup analysis of “Asian” and “PB” under all genetic models . Figure CYP2E1 rs3813867 polymorphism may be associated with the risk of SCC, especially the ESCC cases in Asian populations.We also conducted the overall and subgroup meta-analysis of rs3813867 and SCC risk under the allele (10 case–control studies), carrier (10 case–control studies), homozygote (6 case–control studies), heterozygote (10 case–control studies), dominant (11 case–control studies), and recessive (6 case–control studies) models. The positive results regarding the association between P < 0.05, OR < 1), and the subgroup analysis of “PB” under the allele, carrier, and dominant models . These results suggested a potential link between the c1/c2 or c2/c2 of rs2031920/rs3813867 haplotype and SCC risk, which still requires more case–control studies.The results of overall and subgroup meta-analysis of the rs2031920/rs3813867 haplotype and SCC risk under the allele (five case–control studies), carrier (five studies), homozygote (three studies), heterozygote (five studies), dominant (seven studies), and recessive (three studies) models are shown in Table 2 = 38.3%, P value of heterogeneity = 0.066), however, the random model was utilized for others. The fixed model was used for the allele, carrier, homozygote and recessive models of rs3813867 ; and the allele, carrier, homozygote, dominant, and recessive models of the rs2031920/rs3813867 haplotype .When assessing the heterogeneity level, the fixed model was used for the TT vs. CC model of rs2031920 due to the lack of high heterogeneity  apart from the CT + TT vs. CC model of rs2031920 . Figures Begg’s and Egger’s tests did not provide confirmed evidence of obvious publication bias in the above comparisons  [CYP2E1 rs2031920, rs3813867 SNP and ESCC risk. In this study, we provide evidence that the “T” allele carrier of the rs2031920 polymorphism and the “C” allele carrier of the CYP2E1 rs3813867 polymorphism may be associated with a decreased risk of ESCC, especially in Asian populations because most of the included case–control studies were from China or India., India) . Neverth, India)  and in a, India) . Meta-anCYP2E1 rs2031920 and rs3813867 in the risk of head and neck cancer; they found that the homozygote genotype of CYP2E1 rs2031920/rs3813867 may be linked to the risk of head and neck cancer, especially in Asian populations. Zhuo et al. [CYP2E1 rs2031920/rs3813867 and head and neck cancer risk under the homozygote model. However, the subgroup analysis based of HNSCC was not performed in the two meta-analyses. In our meta-analysis, we failed to observe the statistical relationship between CYP2E1 rs2031920 SNP, rs3813867 SNP, rs2031920/rs3813867 haplotype and HNSCC risk.Tang et al.  selectedo et al.  performeCYP2E1 rs3813867 and the risk of lung cancer in the Chinese population in 2014, and showed a potential link between the “C” allele carriers of CYP2E1 rs3813867 and a decreased risk of lung cancer. In our meta-analysis, very limited data were included after our strict selection; thus, no statistical evidence regarding the role of CYP2E1 rs3813867 in LSCC risk was provided. However, we enrolled five case–control studies [CYP2E1 rs2031920 and found a negative genetic relationship, which was partly in line with the previous data from LSCC subgroup analysis [Cao et al.  selected studies , 33, 39 analysis .CYP2E1 gene was reported [The close linkage disequilibrium between rs2031920 and rs3813867 for the reported –6. For ereported . Howeverreported , 45. ForTo enroll as many eligible case–control studies as possible, a search of five independent online databases  was performed using the overall SCC terms and specific terms, such as ESCC, HNSCC, LSCC and SSCC. Based on our strict criteria, we removed the articles that contained the unconfirmed pathological typing information or failed to provide a genotype frequency distribution in both case and control studies. We observed the absence of large publication bias and the stability of data through Begg’s/Egger’s tests and sensitivity analyses.Despite this, the shortcomings of the small sample size may still have affected our statistical power. Only one case–control study  was inclCYP2E1 gene in our meta-analysis, and we still cannot rule out the potential genetic effect of other CYP2E1 polymorphisms  and the variant combination between CYP2E1 and other related genes .In this study, we focused on the genetic role of two polymorphisms within the P value of heterogeneity = 0.054, I2 = 46.1%) changed to a relatively lower heterogeneity level in the ESCC subgroup . A slight reduction was also observed for the heterozygote model , even though significant between-study heterogeneity existed in the ESCC subgroup. We thus performed another meta-analysis, which only enrolled the available case–control studies of ESCC, and similar results were obtained (data not shown).For rs3813867, we did not observe obvious heterogeneity in the allele, carrier, homozygote and recessive models, only apart from the heterozygote model. Reduced heterogeneity levels were also observed in the ESCC subgroup analysis compared to the overall analysis. For example, in the allele model, a relatively high heterogeneity level in overall meta-analysis (In addition, we observed remarkable heterogeneity for the allele, carrier, heterozygote, dominant and recessive modes of rs2031920. Even though a stable result was detected in the sensitivity analysis, and no decreased heterogeneity level was observed in the subgroup of ESCC compared with overall meta-analysis. This suggested that mixed factors contributed to the source of heterogeneity of specific ESCC subgroups. We tried to analyze the clinical characterizations, such as gender, age or concomitant pathologies, within the enrolled case–control studies. However, in the ESCC, only six eligible case–control studies were included in the ESCC subgroup, and the adjustment data was very limited for categorization. A larger sample size is required to conduct a more in-depth analysis.CYP2E1 rs2031920 and rs3813867 polymorphisms may be associated with the risk of ESCC. However, this conclusion should be confirmed with more extractable case–control studies.In conclusion, our meta-analysis data demonstrated that the


CD44 polymorphisms and cancer risk have been widely investigated previously, whereas results derived from these studies were inconclusive and controversial. We conducted present meta-analysis aiming to explore the association between CD44 polymorphisms and cancer risk. We calculated pooled odds ratios (ORs) corresponding with the 95% confidence intervals (CIs) to make the evaluation clear. Embase, Web of Science, PubMed and Cochrane Library databases were retrieved to identify all eligible publications. As a result, a total of 12 publications comprised 25,777 cases and 27,485 controls fulfilled the inclusion criteria. Nevertheless, the pooled analyses suggested that no significant association was uncovered between CD44  polymorphisms with overall cancer risk. Subsequently, we conducted subgroup analysis for rs13347 polymorphism based on source of control, and we identified a significantly increased cancer risk for the population-based (P-B) group restricted to a recessive model . In conclusion, our meta-analysis demonstrates that CD44 polymorphisms may not represent risk factors for cancer. Future well-designed large-scale case-control studies are warranted to verify our findings.CD44 is one of the commonly recognized stem cell markers, which plays a critical role in many cancer related cellular processes. Relationships between  Malignant tumors pose serious threats to human health and are currently among the top causes of death . In thisCD44 was primarily demonstrated as a receptor for the hyaluronan and lymphocyte-homing receptor . RecentlCD44 were correlated with the risk of many cancers, including BC [et al.'s study [et al. [et al. [CD44 rs353639 polymorphism potentially has a significant effect in BC patients' prognosis. Nevertheless, both rs13347 and rs353639 polymorphisms had no influence on BC risk. Noting these controversial and inconclusive results, we conducted the current meta-analysis in order to determine a more exact relationship between CD44 polymorphisms and the risk of cancer.Several recent studies have demonstrated that many polymorphisms in uding BC , gastricuding BC  and CRC uding BC . In Jian's study , the aut [et al.  verified [et al.  revealedA sum of 12 publications that met the inclusion criteria were enrolled in the quantitative synthesis  and overall cancer risk in all the five genetic models (Table The association between PAdjust < 0.001) rather than hospital-based (H-B) group. Subsequently, we also conducted a subgroup analysis referring to genotyping method. In the MassArray group, statistical heterogeneity preserved significance in all the genetic models . In contrast, no significant association between rs13347 polymorphism and cancer risk was identified for either the RT-PCR or the TaqMan groups  Sensitivity analysis was conducted to evaluate the stability of pooled ORs, in which an individual study will be removed each time in turn from the pooled analyses to detect the influence of individual case-control studies on the pooled ORs. We identified that removal of any single case-control study did not influence the stability of the results. We also generated Egger's funnel plot and conducted Begg's test to assess the publication bias. The shapes of funnel plot appeared symmetrical, indicating no publication bias was existed. These findings were further supported by Egger's funnel plot for the seven Additionally, PRISMA 2009 Checklist for this Meta-analysis was presented in 2≥0.5) between rs187115 and rs353639 polymorphisms. Additionally, LD plots for the YRI population showed a moderate LD value (r2>0.6) between rs11821102 and rs13347 polymorphisms . Subsequently, Sharma et al.[CD44 polymorphisms  and gall bladder cancer (GBC) risk, and they found no significant difference in the frequency distribution of selected polymorphisms in GBC cases when compared with controls at either allelic or genotypic levels in a North Indian population.A previous study performed by Tulsyan et al.  aimed to.'s work , in whicao et al. reports ma et al. re-consiCD44 polymorphisms on overall cancer risk. Thus, our team performed a quantitative meta-analysis to allow for increasing statistical power and provide multiple lines of evidence for the relationship between CD44 polymorphisms and cancer risk. A total of 45 case-control studies were enrolled for the seven polymorphisms . Finally, we identified that the mutated B allele of CD44 polymorphisms was not observed to be associated with an increased risk of cancer. Nevertheless, it is worth noting that our data was not consistent with previously published studies, including a meta-analysis. In a study by Weng et al.[CD44 rs187115 polymorphism with the genotype of at least one G were at an increased risk of developing transitional cell carcinoma (TCC). Around the same time, a similar finding was obtained from a study by Chou et al.[CD44 rs187115 polymorphism may serve as a biomarker for predicting prognosis of late-stage hepatocellular carcinoma (HCC). Furthermore, a study by Xiao et al.[CD44 rs13347 (C > T) polymorphism and NPC development. When the data were stratified based on genotyping method, CD44 rs13347 polymorphism was found to be associated with an increased risk of cancer in the MassArray group in all the five genetic models. Additionally, in the RT-PCR subgroup, we also observed a significant increased association between the rs187115 polymorphism and cancer risk in all the genetic models. Moreover, subgroup analysis based on source of control suggests that a significant association was existed between rs13347 polymorphism and cancer risk in recessive model in P-B group. The existence of this phenomenon may be due to the inconsistencies in control groups. Although most of the controls were chose from healthy populations, many individuals may have suffered from other non-cancer diseases. These differences in control case characteristics could make our findings biased. On the other hand, we also observed significant between-study heterogeneity in our analysis. Absolute meta-regression analysis revealed that the genotyping method introduced substantial heterogeneity. Methodological problems are reflected in the deviations in HWE status, such as the errors in genotyping, the bias of population stratification or selection. Although we did not exclude these studies that were deviated from HWE, we have conducted a subgroup analysis by HWE status. We proved that HWE status did not give rise to bias of results. In addition, the stability of these results were further enhanced by sensitivity analysis.The conclusions from enrolled studies were controversial, and independent studies may not have sufficient statistical strength to precisely identify the effects of ng et al., the autCD44 rs13347 genetic polymorphism and cancer risk, and their ethnicity was restricted to Asians [CD44 and cancer risk, and the ethnicity comprised Asians and Caucasians. The largely increased sample size of current work provides us with more sufficient power to identify some conceal findings. In the end, Shi et al.[CD44 rs13347 (C>T) polymorphism was related to an increased risk of human cancer in Asian people, especially in Chinese populations. Different from their work, we observed the mutated B allele of all CD44 polymorphisms was not associated with the risk of cancer after adjusting. However, several drawbacks in our study should also be noted. Firstly, a relatively small number of studies were enrolled for each polymorphism, with a particularly small number of studies analyzing for the rs353639 polymorphism (only four case-control studies). This limitation may have resulted in an insufficient power for identifying minor association between CD44 polymorphisms and cancer risk. Secondly, further studies are warranted to evaluate the effects of CD44 polymorphisms on cancer risk in different ethnicities. In ethnicity subgroup analysis, the enrolled studies were restricted to Asian and Caucasian populations; data for other ethnicities were not analyzed. Thirdly, the phenotype of our study was a heterogeneous aggregation of a variety of cancer types, and only for part of CD44 polymorphisms, 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 and CC, we were unable to validate the potential effects on these cancers homogeneous or not, which should be investigated in the future. Additionally, several potentially confounding factors were not considered in this study, such as age, sex, smoking and drinking status, (hepatitis B virus) HBV/ (hepatitis C virus) HCV carrier status, environmental factors, and so on.The current meta-analysis comes with some advantages. Firstly, we have conducted a comprehensive search to identify more eligible studies thus, makes our analysis more persuasive and substantial. Secondly, quality of enrolled studies were all assessed by Newcastle-Ottawa Scale (NOS), so low quality studies should be excluded in order to raise the overall quality. Thirdly, subgroup analysis was performed according to cancer type, HWE status and so on at the aim of further deeply exploring the sources of heterogeneity. Fourthly, results were adjusted according to the recognized formula, ensuring the accuracy of the results. In addition, the stability of these studies was further confirmed by sensitivity analysis, and publication bias was tested by Egger's test and Begg's funnel plot. Finally, we have carefully searched for the databases and identified one recent published meta-analysis, which conducted by Shi and his colleagues . They pao Asians . Howeverhi et al. suggesteCD44 polymorphisms might not represent risk factors for cancer. However, our findings require further validation in more well-designed studies with larger sample sizes in order to strengthen our conclusions.Our meta-analysis suggests that CD44 polymorphisms and the risk of cancer. The search terms were as follows: “CD44” AND “SNP OR polymorphism OR mutation OR allele OR variation” AND “cancer OR adenocarcinoma OR carcinoma OR tumor OR neoplasm OR Leukemia OR lymphoma.” The language was restricted to English. These publications were extracted by two reviewers to identify studies specific to various cancers. We then carried out a manual retrieve of the references lists of these enrolled original publications/Reviews to identify additional eligible case-control studies.We carried out a comprehensive literature search on Embase, Cochrane Library and PubMed  to find all relevant publications exploring the relationship between CD44 polymorphisms and cancer risk; 2) they should be case-control/cohort studies; and 3) they should comprise sufficient data . In addition, studies were excluded when they were: 1) case only studies, such as Reviews/comments/case reports and 2) not containing sufficient data.Enrolled studies should meet the following criteria: 1) they should assess the association between Two reviewers (Meng Zhang and Yangyang Wang) performed the data extraction process based on the previously described enrollment criteria. All discrepancies were discussed until consensuses were obtained. In addition, the following characteristics were also extracted from publications: name of first author, publication year, ethnicity of the subjects in the case-control study, source of control, genotype frequency, and etc.CD44 polymorphisms and cancer risk in five genetic models: allele contrast (B vs. A), dominant (BB + BA vs. AA), recessive (BB vs. BA + AA), homozygous (BB vs. AA), and heterozygous (BA vs. AA) models . Subsequently, stratified analyses were performed by cancer type, ethnicity, source of control and genotyping method. We evaluated the statistical heterogeneity assumption by I2 statistics to quantify any inconsistency arising from inter-research variability derived from heterogeneity instead of random chance. An I2 value more than 50% was regarded as significant heterogeneity among these studies. In that case, pooled OR estimations of individual studies were tested by random effect model; if not, fixed effect model will be employed. Moreover, sensitivity analysis was carried out to verify the stability of our results and Egger's regression test and Begg's funnel plot were carried out to evaluate the publication bias. STATA 12.0 software was employed to calculate all the statistical analyses . In addition, Bonferroni corrections were also performed to adjust the results [P<0.05 was regarded as statistically significant. Besides, this study is a systemic review of the literature, so ethical approval was not required.ORs correspondence with 95%CIs were calculated to evaluate the strength of the relationship between  results . P<0.05 http://hapmap.ncbi.nlm.nih.gov/cgi-perl/gbrowse/hapmap24_B36/#search), which comprises CD44 polymorphisms evaluated in the current study. Briefly, populations incorporated in the project including YRI , CHB , JPT  and CEU (Utah residents with northern and western Europe ancestry). Then, Haploview software was employed to conduct the analysis and LD was evaluated by r2 statistics in each of the populations [Data was extracted from the International HapMap Project (ulations .