This study aimed to elucidate the effects of cholecystectomy on the risk of colorectal cancer (CRC) by conducting a meta-analysis of 10 cohort studies.The eligible cohort studies were selected by searching the PubMed and EMBASE databases from their origination to June 30, 2016, as well as by consulting the reference lists of the selected articles. Two authors individually collected the data from the 10 papers. When the data showed marked heterogeneity, we used a random-effects model to estimate the overall pooled risk; otherwise, a fixed effects model was employed.The final analysis included ten cohort studies. According to the Newcastle-Ottawa Scale (NOS), nine papers were considered high quality. After the data of these 9 studies were combined, an increased risk of CRC was found among the individuals who had undergone cholecystectomy (risk ratio (RR) 1.22; 95% confidence interval (CI) 1.08–1.38). In addition, we also found a promising increased risk for colon cancer (CC) , but no relationship between cholecystectomy and rectum cancer (RC)  was observed. Additionally, in the sub-group analysis of the tumor location in the colon, a positive risk for ascending colon cancer (ACC) was found . After combining the ACC, transverse colon cancer (TCC), sigmoid colon cancer (SCC) and descending colon cancer (DCC) patients, we found a positive relationship with cholecystectomy . Furthermore, after combining the ACC and DCC patients, we also found a positive relationship with cholecystectomy  in the sub-group analysis. In an additional sub-group analysis of patients from Western countries, there was a positive relationship between cholecystectomy and the risk of CRC . Furthermore, a positive relationship between female gender and CRC was also found . However, there was no relationship between gender and CC or RC. Furthermore, no publication bias was observed, and the sensitivity analysis indicated stable results.This meta-analysis of 10 cohort studies revealed that cholecystectomy is associated with an increased risk for CRC, CC and ACC, particularly in Western countries. No relationship between cholecystectomy and RC was observed. There was no relationship between gender and either CC or RC, but a positive relationship between female gender and CRC was observed. According to the American Cancer Society, colorectal cancer (CRC) is the third leading cause of morbidity and mortality in the United States (US) . In EuroDue to improvements in living conditions and lifestyle changes, gallbladder diseases (GBDs), such as cholecystitis, cholelithiasis, and gallbladder polyps, are common and costly , with anBoth the biliary tract system and gastrointestinal system are part of the digestive system. Hence, previous studies –16 indicThe cohort studies representing the impact of cholecystectomy on the risk of CRC were searched from the PubMed and EMBASE databases  without restrictions. To identify any additional studies, the reference lists of the relevant articles were also searched. We used the following Entry Terms and Medical Subject Headings (Mesh): (i) "Neoplasms, Colorectal", "Colorectal Neoplasm", "Neoplasm, Colorectal", "Colorectal Tumors", "Colorectal Tumor", "Tumor, Colorectal", "Tumors, Colorectal", "Colorectal Carcinoma", "Carcinoma, Colorectal", "Carcinomas, Colorectal", "Colorectal Carcinomas", " Colorectal Cancer ", " Cancer, Colorectal ", "Cancers, Colorectal", "Colorectal Cancers", OR "Colorectal Neoplasms" [Mesh]; and (ii) ("Cholecystectomies" OR "Cholecystectomy" [Mesh]) OR ; and (iii) "Cohort Study" OR "Cohort Studies" [Mesh].The studies were included in the meta-analysis if they satisfied all of the following criteria: (a) original data source in a cohort study; (b) examined the effects of cholecystectomy on the incidence of CRC, colon cancer (CC) or rectum cancer (RC); (c) the outcome of interest was the incidence of large intestine carcinoma; (d) the exposure of interest was open or laparoscopic cholecystectomy; and (e) the adjusted relative risks (RRs) or standardized incidence ratio (SIR) and the corresponding 95% confidence intervals (CIs) or data to calculate them were provided. All of the reference lists of the retrieved articles were searched by hand to identify any additional studies. Two authors individually examined all of the studies selected from the databases. If the population was studied in more than one study or the data were duplicated, the article with the most comprehensive outcome evaluation or/and the largest sample size was utilized.According to the abovementioned selection criteria, the two authors (HL and YZ) separately evaluated all of the retrieved studies. In addition, a cross-reference search of eligible studies was performed to identify any articles that were not found during the database search. The following information was extracted from each eligible study: the first author’s name; publication year; study region; study period; sample size; number of cases; gender; age; follow-up time; SIR or RR and the 95% CI; and confounding factors. Any disagreements were resolved by discussion or consultation with the co-corresponding authors (LL and ZWT). The Newcastle-Ottawa Scale (NOS) was used to assess the methodological quality of the included studies . The NOS2 test  Figs .By searching the PubMed and EMBASE databases, a meta-analysis was performed to describe the effects of cholecystectomy on the risk of CRC using data derived from only cohort studies. The primary results of our study demonstrated that an increased risk for CRC was found among individuals who had undergone cholecystectomy, which is consistent with previous analyses –16. AddiIt is well known that gene mutations, polyp disease and the long-term effects of anaerobic bacteria (particularly clostridium) on intestinal mucosa play important roles in the pathogenesis of CRC , 43. FurSecondary BAs, particularly lithocholic acid (LCA) and deoxycholic acid (DCA), are regarded as potential tumor-promoting agents in the etiology of CRC  due to tversus males. Moreover, no relationship was observed between cholecystectomy and RC in this study.Additionally, diet and lifestyle also affect the carcinogenesis of CRC . More thThere are several strengths to this study. Cohort studies are able to fully prove causality, which is not possible in observational studies . Therefo2 = 16.0%, P = 0.283). Although only one article [2 = 29.3%, P = 0.167)  . Fourth,= 0.167) . The subIn summary, our meta-analysis proved that cholecystectomy is associated with an increased risk for CRC, CC and ACC, particularly in Western countries. There is no relationship between cholecystectomy and RC. The relationship between gender and CRC, CC or RC is unclear. Further studies on CRC that adhere to strict criteria are needed to further determine the impact of cholecystectomy on the risk of large intestine carcinoma.S1 Table(DOC)Click here for additional data file.


Ezrin, a cytoskeletal protein, is involved in cell adhesion. Several studies have been performed to explore the association between Ezrin and gastrointestinal cancers, but the results are inconclusive. This meta-analysis aims to assess the prognostic value of Ezrin.PubMed and EMBASE were searched. Pooled hazard ratio (HR), odds ratio (OR) and 95% confidence intervals (CI) were utilized to evaluate the association between Ezrin expression and various clinical parameters. 2701 patients from 19 studies were included in the meta-analysis. For gastric cancer, Ezrin expression was associated with tumor grade , TNM stage , lymph node involvement  and overall survival . In colorectal cancer, Ezrin expression was associated with tumor grade , TNM stage , lymph node metastasis , distant metastasis , disease free survival . For esophageal cancer, Ezrin expression was associated with lymph node metastasis  and overall survival .Ezrin expression is significantly associated with tumor grade, TNM stage, and lymph node metastasis in gastric and colorectal cancers. For gastric cancers, Ezrin is useful in predicting distant metastasis. Survival data showed that high Ezrin expression is associated with poor prognosis in gastric, colorectal and esophageal cancers. Our findings suggest that Ezrin might be a potential biomarker in several gastrointestinal cancers. Gastrointestinal cancers include colorectal, gastric and esophageal malignancies. They are very common worldwide, with high mortality rates , 2. EarlEzrin, a cytoskeletal protein, is a member of the ezrin/radixin/moesin (ERM) family. It is involved in a number of signaling pathways that is crucial to cancer progression . For exaA total of 471 studies were retrieved using the strategy described above. 452 studies were excluded after careful screening. The detail of the screening process was shown in Figure In total, 7 studies each yielded data on Ezrin expression in gastric and colorectal cancers, while 5 studies provided information on Ezrin expression and esophageal cancer. The characteristics of the included studies are summarized in Table There were 7 studies with data on Ezrin expression and gastric cancer. Respectively, 6/4/6/3 studies had data regarding Ezrin expression and tumor grade/TNM stage/lymph node involvement/distant metastasis in gastric cancer.P = 0.000). No significant heterogeneity among these 6 studies was observed  , but the heterogeneity among these 4 studies was significant  , with significant heterogeneity among studies  . There, significant heterogeneity among the 3 studies was observed   Figure . Second,) Figure . Third, ) Figure . In addi) Figure .Sensitivity analysis was performed to evaluate the stability of pooled results by sequentially removing individual studies. No single study qualitatively changed the pooled OR. Particularly, 6 out of 7 studies recruited patients with gastric adenocarcinoma. The rest one (Zhang ) includeThere were 7 studies concerning Ezrin expression and colorectal cancer. Respectively, 5/4/3/3 studies provided data on Ezrin expression and tumor grade/TNM stage/lymph node involvement/distant metastasis in colorectal cancer.P = 0.000), where no significant heterogeneity among studies was observed  . But heterogeneity among these studies was significant  , with no significant heterogeneity among studies observed  , where no significant heterogeneity among studies was noted   Figure . Second,) Figure . Third, ) Figure . Also, t) Figure .Sensitivity analysis suggested that the result was reliable. Particularly, 6 out of 7 studies recruited patients with colorectal adenocarcinoma. The rest one (Jorgren ) did notIn total, 5 studies provided data on Ezrin expression and esophageal cancer. Respectively, 4/1/4/1 studies were examined for any association between Ezrin expression and tumor grade/TNM stage/lymph node involvement/distant metastasis in esophageal cancer.P = 0.290), where no significant heterogeneity among studies was observed  . The heterogeneity among these studies was significant   or distant metastasis  in esophageal cancer.The pooled data of 4 studies showed that high Ezrin expression was not significantly associated with tumor grade in esophageal cancer  Figure . The poo) Figure . Only onSensitivity analysis showed that no single study significantly influenced the pooled results .In total, 6 studies were employed to examine possible association between Ezrin expression and survival.P = 0.000), and no significant heterogeneity was observed   for gastric cancer , with no significant heterogeneity observed   in colorectal cancer. The pooled data showed significant association , with no significant heterogeneity noted   patients had confirmed pathological diagnosis of gastrointestinal cancer; (2) Ezrin expression was detected by immunohistochemical methods; (3) the association between Ezrin expression and tumor grade, differentiation, metastasis, overall survival (OS) or disease-free survival (DFS) were evaluated; (4) sufficient data were available for calculation of Hazard ratio (HR), Odds ratio (OR) and 95% confidence interval (95% CI). Duplication of data was avoided by carefully keeping track of the names of all authors and institutions involved in each article. Conference abstracts, case reports, reviews, editorial letters and comments were excluded.Two investigators (Feng Liang and Yangxin Wang) independently extracted data from the relevant articles. Disagreements were resolved through discussion.After strict selection and evaluation, the data were extracted from the included studies as follows: author names, publication years, tumor types, number of patients, Ezrin measurement method, and cut-off used for assessing Ezrin overexpresssion.The Newcastle–Ottawa Scale  was usedet al. [Q test and 2I measurement. P < 0.05 or 2I > 50% indicates significant heterogeneity. Random-effects model is used if the heterogeneity is significant. Otherwise, a fixed-effect model is used. Sensitivity analysis is conducted to help explain any significant heterogeneity among the studies. Egger's and Begg's tests are used to examine publication bias. All statistical analyses are performed using STATA version 11.0 , and all P values are two-sided.Odds ratios (ORs) with 95% confidence interval (CI) are used to evaluate for association between Ezrin expression and several clinical parameters. Hazard ratios (HRs) with 95% CI are utilized to evaluate the association between Ezrin expression and patient prognosis, including overall survival (OS) and disease free survival (DFS). For those where HRs are not given directly, we read Kaplan-Meier curves by Engauge Digitizer version 4.1 and reconstruct the HRs with 95% CI according to the methods described by Tierney et al. . HeterogThis is the first meta-analysis to evaluate the prognostic significance of Ezrin in different gastrointestinal cancers. First, this study indicates that high Ezrin expression is related to tumor grade/TNM stage/lymph node metastasis in patients with gastric/colorectal cancer. Second, Ezrin is capable of predicting distant metastasis in colorectal cancer. Furthermore, high Ezrin expression is associated with poor prognosis according to the pooled data of OS and DFS in gastric/colorectal/esophageal cancers. Therefore, Ezrin may serve as a new biomarker in these cancers to guide clinicians in choosing the most suitable treatment plans. It may also be a potential therapeutic target in the future. However, the limitations mentioned above should be kept in mind. Additional studies with more standardized methodology are needed to confirm our results.


Two recent meta-analyses have been conducted on the relationship between miR-146a polymorphism (rs2910164) and head and neck cancer (HNC) risk. However, they have yielded conflicting results. Hence, the aim of the present study was to conduct a quantitative updated meta-analysis addressing this subject. Eligible studies up to Sep 2016 were retrieved and screened from the bio-databases and then essential data were extracted for data analysis. Next, subgroup analyses on ethnicity, source of controls, sample size, and genotyping method were also carried out. As a result, a total of 9 publications involving 10 independent case-control studies were included. The overall data indicated a significant association between miR-146a rs2910164 polymorphism and HNC risk . Variant alleles of miR-146a rs2910164 may have a correlation with increased HNC risk. Future well-designed studies containing large sample sizes are needed to verify this result. Head and neck carcinoma (HNC) has ranked the sixth most frequent malignancy worldwide, which comprises a number of epithelial cancers originated from oral, nasal cavity, pharynx and larynx . The occThe etiology of HNC remains largely unclear. In recent years, microRNAs (miRNAs) have attracted much attention. MiRNAs are a series of single-stranded short non-coding RNAs that can inhibit gene expression by directly targeting specific mRNAs. They have been suggested to be involved in cellular processes such as cell proliferation, differentiation and apoptosis . Thus, aA popular miRNA, miR-146a, has been suggested to have an association with the development of a variety of disorders . It alsoPotential publications were retrieved from the databases. A total of 29 publications were originally obtained, among which 15 irrelevant papers were firstly excluded. Thus, 14 publications were eligible. Then, 2 review articles , 13 and All publications were written in English, except for one in Chinese . The relThe main results of the meta-analysis are listed in Table For the overall data, a total of ten case-control studies containing 4399 cases and 8777 controls were involved. The pooled ORs for the recessive model  failed to show an association. Nevertheless, borderline increased cancer risks could be shown in both the allelic contrast  and dominant model , implying that variant C allele of miR-146a rs2910164 may have a correlation with increased risk of HNC , suggesting that the publication bias was not evident . Published papers up to Sep 2016 were covered. The following keywords were used: For the literature inclusion, the following criteria were used: (1) papers should concern miR-146a rs2910164 polymorphism and HNC risk; (2) studies should be case-control designed; (3) papers should state adequate information for readers to calculate odds ratios (ORs) and their 95% confidence intervals (CIs). Accordingly, the following criteria were used for exclusion: (1) duplicate publication; (2) papers with insufficient information.The data were extracted by two of the authors independently. If the extracted information was conflicting, a discussion was conducted to reach an agreement. If the agreement could not be reached, another author was consulted and a final decision was made according to the majority of votes. When two or more studies shared the same group of population, only the study including the larger or the largest sample size was selected.The Hardy-Weinberg equilibrium (HWE) was assessed by Fisher's exact test for the controls in each study. The ORs and their 95%CIs were calculated to evaluate the strength of the association between miR-146a rs2910164 polymorphism and HNC risk. The pooled ORs were determined for an allelic contrast model , a dominant model (CC+CG vs. GG), and a recessive model (CC vs. CG+GG).P value for the Q-test greater than 0.05 indicates absence of heterogeneity, and then the ORs were pooled by a fixed-effect model (Mantel-Haenszel) [A chi-squared-based Q-statistic test was performed to assess between-study heterogeneity. A aenszel) ; otherwiaenszel) . The sigaenszel) . To miniaenszel) . All sta


Quantification of the association between menopausal status and risk of lung cancer is inconsistent. We carried out a meta-analysis of available studies to examine this issue.PudMed and Embase databases. Reference lists from selected papers were also reviewed. A random-effect model was used to calculate summary odds ratios (OR) and relative risk (RR) with 95% confidence interval (CI). Publication bias was estimated using Egger regression asymmetry test.Relevant articles were identified by searching Eight eligible studies, including 5 case–control studies and 3 cohort studies, provided data for meta-analysis. Postmenopausal women had a statistically significant increased risk of lung cancer in all included studies  and cohort studies , but not in case–control studies .I2 = 38.40%). To obtain a better indication of relationship, well-designed large prospective studies are required.Overall, there was evidence that postmenopause is related to increased lung cancer risk. However, studies have produced slightly heterogeneous results ( The incidence of lung cancer is increasing in women, in contrast to lung cancer incidence rates in men, which have decreased or stabilized in most countries.,4 Although cigarette smoking is a strong risk factor for lung cancer in women, approximately 10% to 15% of cases occur among never smokers, suggesting other independent factors play an important role in its etiology.Lung cancer is one of the most common cancers in the world and the leading cause of cancer mortality among women in the United States (US).,7 Numerous observational findings have examined the relationship between lung cancer and menopausal types that include premenopause, natural postmenopause, non-natural menopause , and unknown types.,8 However, results have been far from consistent across studies. To further investigate the association between menopausal status and risk of lung cancer, we conducted a meta-analysis of these studies.Numerous recent observational studies of female menopausal type and risk of lung cancer have provided only weak evidence of the role of menopausal type in lung cancer etiology, and results from menopausal studies have been inconsistent.22.1PubMed and Embase databases for relevant published studies before October 2015. No language restrictions were applied, and the following search strategy was used: (“reproductive” or “menopause”) and (“lung” or “pulmonary”) and (“cancer” or “neoplasm” or “carcinoma”). We also perused the references of retrieved articles to identify additional potentially relevant articles.This meta-analysis does not involve patients, and thus does not require institutional review board approval. We conducted a literature search of the 2.2The following criteria were used to include relevant studies for meta-analysis: the study design was an original case–control or cohort study evaluating the relationship between menopausal type (pre- or postmenopause) and lung cancer; the outcome of interest was lung cancer incidence; odds ratio (OR) or relative risk (RR) estimates with 95% confidence interval (CI) were used, with adjustments for potential confounding factors, if applicable. If studies with the same or partially overlapping population were published in more than one study, we considered the most recent or comprehensive study.2.3The following information was extracted from each eligible study: the first author's last name, year of publication, country of study, design, number of cases, menopausal status  of the participants, smoking status, RR or OR, 95% CI, exposure assessment, and adjusted factors. Adjusted and unadjusted (when available) RR estimates  and their variance  were extracted. ORs were treated as RRs. We extracted the ORs and RRs that were adjusted for the greatest number of potential confounders.In the 8 studies women were asked their menopausal status, for example, whether they were pre- or postmenopause, age at menopause, and type of menopause . For surgical menopause, they were asked whether they had a hysterectomy with or without oophorectomy and number of ovaries removed. Participants were also asked their smoking status , and their environmental tobacco smoke exposure at home (during childhood and adulthood) and in the workplace. To reduce reporting bias, the studies did not reveal interviewer and participant interests, or inform the status of the participants to the interviewers.2.4 and CIs to give the RR and CI of a comparison different from those provided. We used this method to estimate the relationship between subjects and exposure categories. The RR and CI for the specified contrast are calculated by combining the estimated numbers of subjects into baseline  and comparison groups. The program also estimates heterogeneity and trend coefficients among the selected categories, and can be used for case–control and prospective studies.Studies gave RR and CI for at least 3 exposure categories , each compared with a single unexposed group (postmenopause). However, we needed the overall RR and CI for exposure versus no exposure. Relative risk estimation (RREst) is a spreadsheet program that manipulates nonindependent RRs2.5 Statistical heterogeneity among studies was assessed using Cochrane Q test at the P < .05 level of significance and the I2 statistic. Subgroups were analyzed by study design and geographical area. We also performed a sensitivity analysis by omitting one study before pooling study-specific RRs to examine the influence of individual studies. We used Begg funnel plots and Egger linear regression test to assess publication bias.,13 Statistical analyses were done using STATA Statistical Software, version 12.0 .We used the DerSimonian and Laird random-effects model to estimate an RR summary considering both within- and between-study variation.33.1PubMed and Embase databases and identified 4406 articles . A total of 271 duplicated articles and 4089 articles were removed at first screening of titles and abstracts. Full-text copies of the remaining 46 potentially eligible publications were obtained. Of these, 38 articles were excluded because they did not meet the following criteria: result was not lung cancer incidence (n = 3), no RR or OR for menopausal status (n = 32), and review (n = 3). The remaining 8 studies were included for meta-analysis ,,19 Germany (n = 1), Japan (n = 2),,20 China (n = 1), Canada (n = 1), and Italy (n = 1). Total sample size of the cohort studies was 383,391 subjects with 1197 lung cancer events; the total lung cancer cases for the case–control studies were 2411, and control cases in the case–control studies totaled 4499.Table 3.3We used the RREst program to enter each category description, and values/estimates for number of subjects, RR, and lower/upper 95% CI categories, to obtain specified contrast results. As an example, data for Wu et al included natural menopause  and surgical menopause  for which the program provided new data . Using the same process provided data for Kreuzer et al , Liu et al , Pesatori et al , and Gallagher et al . These results were combined with the data for Matsuo et al, Koushik et al, and Weiss et al for the STATA statistical analyses.3.4P = .123, I2 = 38.4%). A statistically significant 44% increased lung cancer risk was found to be associated with postmenopausal women. When stratified by study design, we found that postmenopause was associated with a statistically significant 39% increased lung cancer risk in cohort studies , but not in case–control studies .Overall risk estimates for lung cancer were elevated among postmenopausal women compared with premenopausal women .We performed subgroup analyses by study design, study geographical area, assessment of exposure, adjusted for confounders, smoking, age, and smoking and age Table . Postmen4I2 = 63.50%); however, results from the 3 cohort studies were null. Case–control studies are more susceptible to recall and selection bias than cohort studies and the results from cohort studies may be more reliable. In subgroup analysis by study location, we found that postmenopause was associated with increased lung cancer risk in the US and Europe, but not in Asia, suggesting that ethnic genetic background and lifestyle differences should be considered. When we stratified the analysis by assessment of exposure, a significant association was found among studies that used interviews but not among studies that used mailed questionnaires. This contrast may be due to the differing techniques or chance alone.As far as we know, this is the first meta-analysis evaluating the relationship between menopausal status and lung cancer risk. In the extensive search, only 8 studies  that met the inclusion criteria, although the total number of subjects was considerable . Control study analysis suggested that postmenopausal women were at a higher risk of developing lung cancer than premenopausal women, but we failed to demonstrate this with the case–control studies. There was statistically significant heterogeneity of the 5 case–control studies (There are several potential limitations. First, primary data were unavailable for most of the studies and additional adjustments for potentially important covariates could not be performed. Second, studies used different scales to evaluate menopausal status, so we could not analyze the risk of lung cancer in women who underwent natural or surgical menopause compared with those who had not undergone menopause. Differences in statistically adjusted RRs across studies may account for some unexplained heterogeneity. Third, the result may be influenced by age and smoking because cigarette smoke exposure is already established as an independent risk factor for developing lung cancer, but when we adjust for smoking and age, the result has not changed. Finally, we cannot preclude the possibility that other unpublished studies may have been missed despite extensive literature search in our meta-analysis. The potential publication bias may have been because studies with null effects were less easily published than those with positive effects, which made it different for us to obtain, although Egger test or Begg test did not reveal the present of publication bias.A potential confounder is the inability to determine the role of natural or surgical menopause in lung cancer, as these data were incomplete. Despite its limitations, the study represents a possible connection between menopausal factors and lung cancer, providing direction for further studies of the role of menopause in the diagnosis, treatment, and survival of women with lung cancer.In conclusion, current data indicate that menopausal status may increase the risk of lung cancer in women. The mechanisms involved are likely to be complex. Additional studies are warranted to extend our findings and to clarify the unknown mechanisms. Further studies are needed to investigate the effect of natural and surgical menopause on lung cancer risk and which type of lung cancer. The number of epidemiological studies was small in our meta-analysis; further well-designed, large prospective studies are needed to elucidate this relationship.


Each year, 430,000 people are diagnosed with bladder cancer. Due to the high recurrence rate of the disease, primary prevention is paramount. Therefore, we reviewed all meta-analyses on modifiable risk factors of primary bladder cancer. PubMed, Embase and Cochrane database were systematically searched for meta-analyses on modifiable risk factors published between 1995 and 2015. When appropriate, meta-analyses (MA) were combined in meta–meta-analysis (MMA). If not, the most comprehensive MA was selected based on the number of primary studies included. Probability of causation was calculated for individual factors and a subset of lifestyle factors combined. Of 1496 articles identified, 5 were combined in MMA and 21 were most comprehensive on a single risk factor. Statistically significant associations were found for current (RR 3.14) or former (RR 1.83) cigarette smoking, pipe (RR 1.9) or cigar (RR 2.3) smoking, antioxidant supplementation (RR 1.52), obesity (RR 1.10), higher physical activity levels (RR 0.86), higher body levels of selenium (RR 0.61) and vitamin D (RR 0.75), and higher intakes of: processed meat (RR 1.22), vitamin A (RR 0.82), vitamin E (RR 0.82), folate (RR 0.84), fruit (RR 0.77), vegetables (RR 0.83), citrus fruit (RR 0.85), and cruciferous vegetables (RR 0.84). Finally, three occupations with the highest risk were tobacco workers (RR 1.72), dye workers (RR 1.58), and chimney sweeps (RR 1.53). The probability of causation for individual factors ranged from 4 to 68 %. The combined probability of causation was 81.8 %. Modification of lifestyle and occupational exposures can considerably reduce the bladder cancer burden. While smoking remains one of the key risk factors, also several diet-related and occupational factors are very relevant. The International Agency for Research on Cancer (IARC) estimated that worldwide 430,000 people were diagnosed with bladder cancer in 2012, with an age-standardised rate (ASR) of 5.3 per 100,000 people. Incidence of bladder cancer is particularly high in males with 77 % of the cases, ranking it the 7th highest on incidence and 9th highest on mortality. For females it is the 19th highest on incidence and 17th highest on mortality. The incidence of bladder cancer is almost 3 times higher in more developed countries compared to less developed countries  , “urinary bladder neoplasms”[All Fields], “bladder cancer”, combined with systematic[sb], meta-analysis[Title/Abstract], meta-analysis[Publication Type], systematic review[Text Word], pooled analysis[Text Word].A first screening was done based on the title and abstract. Articles were considered for inclusion if they were MAs or pooled analyses describing the association between modifiable risk factors and development of bladder cancer, written in English. Inclusion was limited to articles published from 1995, as during the 1990s systematic reviews and meta-analyses started to replace narrative reviews, and methods and terminology were well established by that time . When thThe exposure of interest was any modifiable risk factor. ‘Modifiable’ was defined as a risk factor that is in any way reasonably modifiable, meaning that genetic and medical conditions were excluded as these are generally not modifiable. The outcome of interest was defined as incidence of bladder cancer as a binary outcome. Studies that included estimates based on prevalence or mortality were used as a proxy for incidence when estimates on incidence only were not available. Studies presenting results on genitourinary cancers without giving estimates for bladder cancer alone and studies reporting on urothelial cancer alone were excluded. Additionally, pooled risk estimates had to be presented with the accompanying 95 % confidence interval, accepted as odds ratio (OR), risk ratio (RR), hazard ratio (HR), standardised incidence ratio (SIR), or effect size (ES) if the latter consisted of a combination of the previously mentioned ratios. For the purpose of this paper, these estimates are all referred to as relative risks (RR).Data was extracted by AA and all entries were double-checked completely for errors by KS. Any disagreement was discussed between the review authors until consensus was reached. When needed, authors of publications were contacted for additional information.For each article, the following data was extracted when available: first author, year of publication, journal, risk factor(s) studied, number of events, types of studies included, number of studies included, type of risk estimate , outcome variable , and per MA : the exposures that were compared, the risk estimate and the 95 % confidence interval, a measure of heterogeneity, the number of primary studies included, and a measure of publication bias.The estimates to be included had to be based on a minimum of two primary studies. They were selected based on the following list of decisions, and in this order: (1) bladder cancer only over urothelial cancers, (2) overall estimates over sub-group estimates, (3) more adjusted estimates over less adjusted estimates, (4) smoking adjusted over other adjustments, (5) incidence over prevalence or mortality, (6) random effects model over fixed effects model, and (7) pooling of cohort studies over pooling of case–control or cross-sectional studies.If only one article was available on a specific risk factor, these estimates were included directly. If multiple articles were available, we selected the most comprehensive MA. This could occur in one of two ways. (1) The preferred method was to combine two MAs in a MMA to obtain an overall estimate based on the largest number of studies. However, we considered this only appropriate if the level of overlap between the primary studies included in each publication was no more than 50 %, measured as the percentage of studies of the smaller publication that overlapped with those of the larger publication. In addition, the exposure categories had to be comparable enough to be combined. (2) If, by following this rule, it was not possible to combine publications, the most comprehensive publication was included, based on the largest number of included primary studies. Basing this on the largest number of studies, rather than the largest sample size, was done because it was often not possible to determine the publication with the largest sample size due to insufficient data in the publication.2 statistic [p values of 0.05 or below, or a confidence level of 95 %, were considered statistically significant.MMA was performed using a random-effects model. Heterogeneity was assessed between the included MAs using the Itatistic . All staIn order to provide a measure of public health impact, a probability of causation  (POC) waPOC can be interpreted as the percentage of exposed cases that are attributable to the exposure (e.g. the number of cases among smokers that are attributable to smoking). In case of a protective factor this should be read as the proportion of non-exposed cases that are attributable to the fact that they are non-exposed (e.g. the number of cases among people with low fruit and vegetable intake that are attributable to the low fruit and vegetable intake).The funding body was not involved in the study design; in the collection, analysis, or interpretation of data; in the writing of the report; or in the decision to submit the paper for publication.Figure Four general topics emerged from the literature search with regard to lifestyle: dietary factors, consumption of tobacco, alcohol consumption, and physical activity.2 = 58.4 %; n = 10).The MA by Yao et al.  was sele2 = 54.9 %; n = 27) as were citrus fruit . Yao et al. [MA by Yao et al.  was seleo et al.  reported2 = 0.0 %; n = 21 + 8) and cruciferous vegetables specifically , resulted in a reduced risk of bladder cancer.By pooling results from Yao et al.  and Stei2 = 82.8 %; n = 14).Bai et al.  was the 2 = 33.1 %; n = 7) but not for overall and green tea consumption.The MA by Qin et al.  was sele2 = 94.2 %; n = 18 + 9 + 6).Six MAs –28 reporOnly one MA by Boyle et al.  was iden2 = 70.1 %; n = 16; skim milk: RR 0.47; 95 % CI 0.18–0.79; I2 = 0; n = 2; fermented milk: RR 0.69; 95 % CI 0.47–0.91; I2 = 62.5; n = 5). In contrast, consumption of whole milk was associated with a significantly increased risk of bladder cancer .As reported by Mao et al. , which wThe MA by Li et al.  was the The MA by Li et al.  was seleThe MA by Pelucchi et al.  was consThe MA by Li et al.  was the 2 = 46.3 %; n = 11). When looking only at intake of vitamin A, the association was significantly protective for vitamin A supplementation, consisting of beta-carotene, and/or retinol supplementation  but not for dietary intake .The MA by Tang et al.  was sele2 = 0.0 %; n = 2).Jeon et al.  investig2 = 43.7 %; n = 20).The MA by Wang et al.  was the The MA by Liao et al.  investig2 = 0 %; n = 15).The MA by Wang et al.  was sele2 = 28.9 %; n = 13). The association remained significantly reduced when pooling studies on dietary folate only , but not on supplemental folate only .The association between folate intake and risk of bladder cancer was studied only in a MA by He et al. . They foTwo MAs , 43 wereSelenium supplementation was studied in one MA by Vinceti et al. . They me2 = 0.0 %; n = 4).One MA by Myung et al.  reportedThe association between dietary acrylamide intake and bladder cancer was studied in one MA, authored by Pelucchi et al.  They rep2 = 8.8 %; n = 12) but not among pre-obese individuals .Two MAs , 47 repoEight MAs , 48–53 wVan Osch et al.  reportedHemelt et al.  also repTwo MAs , 53 addi2 = 35.6 %; n = 8) compared to never-smoking individuals with no exposure to passive smoking.Van Hemelrijck et al.  reportedUse of smokeless tobacco, including chewing tobacco, oral snuff, and unspecified smokeless tobacco, was studied in two MAs , 55. LeeKeimling et al.  wrote thTurati et al.  updated Multiple MAs were identified on the association between occupational factors and bladder cancer. Most MAs reported on a single occupation or occupational group, or a single specific occupational exposure Appendix . MoreoveThe estimates for probability of causation (POC), displayed in percentages, are presented in Table A POC was calculated for four factors combined, which were considered sufficiently independent: total fruit and vegetable consumption, processed meat consumption, smoking, and physical activity. To avoid overlap, the only estimate included for smoking was for cigarette smoking as most smokers are  cigarette smokers (78.7 %) and only 4.5 % was found to be pure pipe and/or cigar smoker . The comIn total, 12 lifestyle factors and 48 occupational factors emerged that significantly increased or decreased the risk of bladder cancer. Such numerous modifiable risk factors present an opportunity and great potential for prevention. Below, the significantly associated risk factors will be discussed.The most important risk factor proved to be smoking, particularly the risk was highest for current smokers of cigarettes (RR 3.14),  but alsoCharacteristics of the tobacco smoked may influence this association. In previous studies, individuals smoking black tobacco had a higher risk compared to smokers of blond tobacco, due to the higher concentrations of carcinogens in black tobacco –71. AlsoTobacco smoke contains numerous carcinogens that contribute to the initiation and promotion of tumour development. These chemicals are renally excreted, making it directly toxic to the human urinary bladder. During metabolisation of these compounds, DNA-adducts are formed, leading to permanent genetic mutations. If this occurs in an oncogene or tumour suppressor gene, it may result in uncontrolled growth and eventually cancer.Genetic mutations, such as NAT2 slow acetylation and GSTM null genotypes, have previously been shown to be associated with increased susceptibility to cancer . The enzBecause the metabolites of many food groups are excreted by the urinary tract, dietary factors have often been suggested and researched as risk factors of bladder cancer . SeveralOne key group of foods that was associated with a lower risk of bladder cancer were fruits and vegetables. Besides these groups as a whole, intake of both citrus fruit and cruciferous vegetables were researched. The effect is suggested to be due to the vitamins, minerals, and phytochemicals that fruit and vegetables contain, which have antioxidant properties , 85. AntIt should be noted that significant publication bias was found in the MA on overall and citrus fruit consumption . BecauseThe MAs included in our review that studied vitamins separately, came to supporting conclusions. For all but vitamin C, higher intakes were associated with a significantly reduced risk of bladder cancer, with effects on −39 % for selenium, −18 % for vitamin A and E, and −16 % for folate.Vitamin A intake was found to be negatively associated with bladder cancer when data on dietary intake, supplementation, and blood levels were combined. However, when studied by intake source separately, the protective effect was only found for vitamin A supplementation but not for dietary intake . FurtherVitamin E consists of a group of lipid soluble molecules, including four tocopherols and four tocotrienols. Specifically γ-tocopherol, δ-tocopherol, and tocotrienols inhibit several inflammatory pathways,  which isAlthough higher levels of selenium protected against bladder cancer, causation could not be confirmed through supplementation intervention , The speHowever, despite these suggested protective properties of antioxidants, analysis of multiple types of antioxidant supplementation together, showed a 52 % increased risk of bladder cancer . An explBecause consumption of processed meat but not red meat was associated with bladder cancer, the effect may result from different forms of mechanical, chemical, and enzymatic treatment that processed meat has been exposed to. One hypothesis suggests that nitrite, which is used as a colour and flavour preservative in processed meat, combines with secondary amines from proteins to form nitrosamines that are found to be carcinogenic to, amongst other organs, the bladder . AlthougThe protective effect of folate, a vitamin predominantly found in vegetables, is suggested to be through its role in DNA synthesis, repair, and methylation . In prevIn line with previous findings for colorectal cancer  and breaPrevious reviews have also mentioned other dietary factors increasing the risk of bladder cancer, among which higher intake of soy, fat, barbecued meats, and artificial sweeteners , 104. HoWhereas increased intake of fluids dilutes urine and increases micturition, leading to a reduced exposure of the bladder to carcinogens, it may also increase exposure when the fluid itself contains carcinogens . For exaPhysical activity was found to have a small but significant protective effect against bladder cancer. The effect found was not modified by BMI or smoking, as the authors explain in their MA . AlthougOccupational factors are considered the second most important risk factor for bladder cancer after smoking , 110. OvParticularly workers exposed to aromatic amines, PAH, tobacco and tobacco smoke, combustion products, and heavy metals are at an increased risk. However, these findings will not be further discussed and repeated here as the MA by Cumberbatch et al.  offers aInterestingly, farmers, gardeners, teachers, and forestry workers were found to be at a significantly lower risk. Farmers have been studied numerous times with regard to disease risk and have been suggested to be at both an increased and decreased risk of bladder cancer. It is thought of as a risk factor due to the exposure to chemicals such as pesticides, viruses, and other exposures. Although an increased risk was found in association with exposure to some pesticides, the evidence is not conclusive . Both thFinally, in addition to exposure to carcinogens on an occupational basis, an increased risk of bladder cancer in certain occupational groups may also be the result of fewer possibilities for micturition during working hours. This may apply to, for example, sales workers, professional drivers, or other occupations where bathroom visits are limited to break times.The calculated POCs show that by adopting the right lifestyle, while considering harmful environmental and occupational exposures, a large proportion of the burden of bladder cancer could be prevented. Particularly for smoking the POC was high, highlighting that smoking cessation and prevention of smoking initiation should remain high on the agenda. In addition, a diet rich in fruit and vegetables, particularly citrus fruit and cruciferous vegetables, and low in processed meat should be stimulated. Particularly working conditions involving exposure to aromatic amines, PAHs, tobacco and tobacco smoke, combustion products, and heavy metals should be given priority to reduce exposure to these compounds.The quality of a MA is strongly determined by the quality of the primary studies included. Not all studies corrected for smoking status, which is the most important risk factor for bladder cancer. The effects of this may be large, leading to an under- or overestimation of the true effect size. To minimise the effect of confounding in our study, only the most adjusted estimates from each MA were selected. Selection of the most comprehensive publication was based on the largest number of primary studies included, rather than the largest sample size. The latter would have been more desirable as the power of a study depends on it, but this data was often unavailable in the publications and number of studies included had to be used as a proxy. Results of MMA could be affected by duplicate inclusion of primary studies, leading to inflated precision and homogeneity . TherefoThe main limitation with regard to the combined POC is assumption of independence of risk factors. It is well known that lifestyle behaviours tend to cluster,  indicatiIn this study, estimates of bladder cancer incidence were given priority over estimates of prevalence or mortality. Although the majority of included studies specified the outcome measure that was included or provided stratified results, not all did Table . BecauseHeterogeneity may always be an issue in MAs. Several of the studies included in the current review show moderate to high levels of heterogeneity. Whereas some identified study design, gender, or geographical region as possible source of heterogeneity , 22, 31,It should be noted that different classification criteria may be applied in studies to determine cases, including the TNM classification and stage grouping. Because any effect would be unlikely and small, this potential difference was not taken into consideration when selecting estimates.This publication is a review of MAs. Additional risk factors for which no MA is available may prove to be of importance, but have not been included in this review. This study will help to identify gaps in knowledge with regard to MAs available.It was our responsibility to give an objective and complete overview of the literature available. By using a protocol, including only MAs and by dealing appropriately with unexpected issues, we believe to have provided an overview of the risk factors associated with bladder cancer with high levels of objectivity and reliability.In conclusion, the burden of bladder cancer could be significantly reduced through modification of lifestyle, environment, and occupational exposures. In fact, having numerous modifiable risk factors, among which multiple with a substantial RR, makes bladder cancer one of the most preventable diseases. Although substantial risks were found for, for example, smoking, some identified risk factors ‘only’ led to an increase or decrease of 10–15 %. This could be considered a minor effect, but reducing prevalence of a number of smaller risk factors together, could still result in significant overall risk reductions in the general population. Theoretically the included risk factors are, at least partially, modifiable. However, actually changing them is not always easily achieved, neither for one individual, let alone on a population level. Therefore, it is important that preventative and protective strategies pay attention to behaviour change in order to capture the greatest potential of risk reduction. When change is achieved, impact will not remain restricted to an improvement in the incidence of bladder cancer, as most of the included risk factors also play a role the risk of multiple other chronic diseases such as cancer or cardiovascular diseases, –118 and


After retrieving articles from online databases, eligible studies were enrolled according to the selection criteria. Stata/SE 12.0 software was utilized to perform the statistical analysis.Passociation<0.05). However, no significant difference between liver cirrhosis cases and the control group was observed for HFE C282Y and H63D . In addition, we found that HFE C282Y was statistically associated with increased HCC susceptibility in the overall population, while H63D increased the odds of developing non-cirrhotic HCC in the African population . Moreover, a positive association between compound heterozygosity for C282Y/H63D and the risk of NAFLD and HCC, but not liver cirrhosis, was observed.In total, 43 articles with 5,758 cases and 14,741 controls were selected. Compared with the control group, a significantly increased risk of NAFLD was observed for the C282Y polymorphism in the Caucasian population under all genetic models and for the H63D polymorphism under the allele, heterozygote and dominant models (all OR>1, Our meta-analysis provides evidence that the HFE C282Y and H63D polymorphisms confer increased genetic susceptibility to NAFLD and HCC but not liver cirrhosis. Additional well-powered studies are required to confirm our conclusion. Hepatocellular carcinoma (HCC) often occurs as the end-stage or more aggressive form of many progressive chronic liver diseases, such as NAFLD , liver cirrhosis, and chronic viral hepatitis –4. NAFLDHuman hereditary hemochromatosis (HFE) gene, first identified by Feder JN et al. in 1996, is located on the short arm of chromosome 6 6p21.3) . The HFE.3 . The Although several previous meta-analyses on the association of HFE genetic variants and NAFLD and HCC risk have been reported , 23–26, The current meta-analysis followed the guidelines  of “Prefst, 2016), including PubMed, EMBASE, Web of Science (WOS), Scopus and China National Knowledge Infrastructure (CNKI), were thoroughly searched for potential records. Five electronic databases  test in the control group. For unavailable or missing data, we attempted to contact the corresponding or first author through E-mail or the ResearchGate website.Potential records from the systematic literature search were selected for eligible studies. We utilized EndNote X7 software to remove duplicate records and screened the title and abstract to exclude unrelated records based on the following criteria: (1) Review or book; (2) Not clinical data; (3) Other gene; (4) Other disease; (5) Case, trial, or non-polymorphism; (6) Meta-analysis; and (7) Meeting abstract. After the assessment of eligibility, all eligible articles were required to provide sufficient data regarding the genotype distribution of HFE polymorphism in both the case and control groups. The following basic information was extracted independently by the authors (QY BXQ WLY FMW): name of the first author, year of publication, country, ethnicity, sample sizes, source of control, genotyping methods for SNP, disease definition, diagnostic method, genotype frequencies, and the http://www.ohri.ca/programs/clinicalepidemiology/oxford.html [Three authors (QY BXQ WLY) independently assessed the methodological quality of the included studies, according to the Newcastle-Ottawa Scale (NOS) system, which is available from ord.html . The NOSPassociation were determined through Mantel-Haenszel statistics using Stata/SE 12.0  software. A Passociation<0.05 indicated a significant difference between the case and control groups. Five inheritance models, namely, allele , homozygote , heterozygote , dominant  and recessive models , were applied.The values of pooled odds ratios (ORs), 95% confidence intervals (CIs) and 2 test were performed to evaluate the potential heterogeneities among studies. A random-effect model was used when the existence of significant heterogeneity was not excluded . To investigate the potential sources of heterogeneity, subgroup analyses based on ethnicity, source of controls, genotyping methods, HWE and disease type  were conducted. In addition, both Begg’s test and Egger’s test were performed to assess the potential publication bias, and sensitivity analysis was conducted to evaluate whether the results were statistically stable. The PHWE value was obtained from a chi-squared test, and PHWE greater than 0.05 was considered as being in agreement with HWE.Cochran’s Q statistic and IThere were 43 eligible articles with 5,758 cases and 14,741 controls included in our meta-analysis , 29–67, A total of 1,287 potential records were obtained from the PubMed (n = 391), EMBASE (n = 219), WOS (n = 583), Scopus (n = 90) and CNKI (n = 4) databases, and 469 duplicate records were removed by EndNote software. Additionally, 699 records were excluded by screening the title and abstract for the following: Review or book (n = 317), Not clinical data (n = 34), Other gene (n = 48), Other disease (n = 166); Case, trial, or non-polymorphism (n = 114); and Meta-analysis (n = 20). The authors (QY BXQ WLY FMW) independently extracted the data from 119 full-text articles and removed 76 articles, including 24 articles, which were meeting abstracts, and 52 articles, which lacked usable data. A flow diagram of the literature search strategy for the meta-analysis is presented in 2 >25%, Pheterogeneity<0.1). The pooled results in Passociation = 0.012), heterozygote  and dominant models  but not in the other models.We first investigated the genetic association between the HFE C282Y polymorphism and the susceptibility to NAFLD. A random-effect model was used for Mantel-Haenszel statistics due to the high degree of heterogeneity (PHWE>0.05 and PHWE <0.05) and specific disease type (NASH). As shown in PHWE >0.05 subgroup . These data suggested that HFE C282Y may be linked to the risk of NAFLD, especially in the Caucasian population.Moreover, subgroup analyses under all genetic models were conducted based on ethnicity , source of controls (PB and HB), genotyping methods (PCR-RFLP and other), HWE . As shown in Passociation = 0.003), HD vs HH , and HD+DD vs HH . A similar significant difference was observed in the subgroup analysis for the Asian population, PB, PCR-RFLP, and NASH . Therefore, the HD genotype of HFE H63D contributes to increased NAFLD susceptibility.An association between HFE H63D and NAFLD risk was also detected. No large heterogeneity was detected . No significant difference between the control and case group was observed for C282Y under all genetic models (Passociation>0.05). For H63D, as shown in 2 = 0.0%, Pheterogeneity>0.1), whereas a random-effect model was used for the others . There were no significant differences under the majority of comparisons in the meta-analysis and subsequent subgroup analysis (Passociation>0.05), except for the allele, homozygote and recessive models in the Asian population and the allele model in the PHWE>0.05 group. These data failed to provide strong evidence of a significant correlation between HFE C282Y and H63D and liver cirrhosis risk.The data in 2>25%, Pheterogeneity<0.1), and an increased HCC risk was observed under all genetic models in the overall population . A similar significant difference was observed in the pooled analysis in the PB, PHWE>0.05 and cirrhosis (-) subgroups under the allele, homozygote, dominant and recessive models (Passociation<0.05). For H63D polymorphism, no increased HCC risk was observed in the overall population (Passociation>0.05); however, there was a significant difference in the African population and the cirrhosis (-) subgroup (Passociation<0.05). Our data demonstrated that HFE C282Y may increase the odds of developing HCC, while HFE H63D is more likely associated with susceptibility to HCC without cirrhosis in the African population.A random-effect model was used to analyze the genetic association between HFE C282Y and HCC risk (Passociation = 0.023) and for HCC in the overall population  but not for liver cirrhosis . These data suggested that the effect of C282Y+H63D compound heterozygosity may contribute to an increased risk of NAFLD and HCC.Next, to study the potential role of compound heterozygosity for C282Y/H63D in susceptibility to NAFLD, liver cirrhosis and HCC, we performed a meta-analysis and subgroup analysis. As shown in PBegg>0.05, PEegger>0.05). For H63D polymorphism, small publication bias was observed in the DD vs HH  and DD vs HH+HD models . However, there was no evidence of publication bias in the other models . No obvious publication bias was observed for compound heterozygosity for C282Y/H63D . Moreover, the sensitivity analysis further confirmed the statistical stability of our results  , 68–70. Passociation value, OR and 95% CI were calculated using Mantel-Haenszel statistics under the allele, homozygote, heterozygote, dominant and recessive models. (3) NAFLD is not a uniform clinical disease entity [In 2007, Ellervik C and colleagues reported that the HFE C282Y mutation might be linked to an increased risk of NASH under a homozygote meta-analysis model . In 2011e entity , 71, andProgressive iron overload in the liver was considered a key factor for the presence of liver injury, chronic inflammation, fibrosis, cirrhosis, liver failure and cancer , 73. TheDiverse conclusions on the role of HFE polymorphism in HCC risk have also been obtained. For instance, the C282Y heterozygous genotype was reported to be associated with susceptibility to HCC . H63D waThe disadvantages of our systematic review and meta-analysis are as follows. (1) The eligible articles contain relatively small sample sizes. For instance, only two case-control studies were enrolled in the meta-analysis of the Asian subgroup. (2) Four moderate-quality studies were included in our meta-analysis , 55, 57.In summary, our updated systematic review and meta-analysis confirmed the role of HFE C282Y in an increased HCC risk and provided new evidence that H63D is more likely to be associated with susceptibility to non-cirrhotic HCC in the African population. A significant correlation between HFE C282Y and H63D polymorphism and NAFLD susceptibility was obtained. Furthermore, we found that the HFE mutations failed to increase the odds of developing liver cirrhosis. The first evidence regarding the positive genetic relationship between compound heterozygosity for C282Y/H63D and the risks of NAFLD and HCC was demonstrated.S1 Fig(A) Forest plot analysis; (B) Begg’s test; (C) Egger’s test; (D) Sensitivity analysis.(TIF)Click here for additional data file.S2 Fig(A) Forest plot analysis; (B) Begg’s test; (C) Egger’s test; (D) Sensitivity analysis.(TIF)Click here for additional data file.S3 Fig(A) Forest plot analysis; (B) Begg’s test; (C) Egger’s test; (D) Sensitivity analysis.(TIF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(DOCX)Click here for additional data file.S4 Table(DOCX)Click here for additional data file.S5 Table(DOCX)Click here for additional data file.S6 Table(DOCX)Click here for additional data file.S7 Table(DOCX)Click here for additional data file.S8 Table(DOCX)Click here for additional data file.S9 Table(DOCX)Click here for additional data file.


Here, we performed a meta-analysis to evaluate whether promoter hypermethylation of the MGMT gene contributed to gastric pathogenesis.The association of MGMT  databases. The Newcastle–Ottawa Scale was applied to assess methodological quality of the included studies. Pooled odds ratios (ORs) and 95% confidence intervals (95% CIs) were calculated to evaluate the association of MGMT promoter hypermethylation with gastric pathogenesis. Moreover, STATA 12.0 software was used to summarize the extracted data in this meta-analysis.MGMT promoter hypermethylation in the GC group (32.97%) was significantly higher than those in the control group (18.00%) . When stratified by cancer subtype, the results indicated that the frequency of MGMT promoter hypermethylation was significantly higher in gastric adenocarcinoma than in control group . In addition, MGMT promoter hypermethylation significantly promoted distant metastasis and lymph node (LN) metastasis of gastric tumor . A significant association between MGMT promoter hypermethylation and TNM-stage was also found in the present meta-analysis .Seventeen studies, comprising 1736 cases and 1291 controls, were included in this meta-analysis. The frequency of MGMT gene-promoter hypermethylation was significantly associated with an increased risk of GC, especially in Asians. Furthermore, MGMT gene-promoter hypermethylation might be correlated with the distant metastasis and LN metastasis of GC.The results of this meta-analysis suggested that  Although advances of early detection have led to a decline in incidence rates of GC, a lot of GC patients were still diagnosed at late stage, and patients usually died of metastasis. Many studies have confirmed that the development of GC was a multifactorial process that involved some environmental factors, multiple genetic, and epigenetic alterations such as Helicobacter pylori (H. pylori) infection, high salt intake, smoking, gene variation, and gene methylation.,5 In terms of risk evaluation, H. pylori infection was significantly associated with the risk of GC. However, H. pylori infection was insufficient for predicting the risk of GC. Thus, other biomarkers were still needed to be identified to improve our understanding of gastric carcinogenesis. Epigenetic changes such as promoter methylation and histone acetylation played an important role in cancer development. A number of tumor-related genes were often silenced by those deregulated modifier that produced epigenetic changes. Most deoxyribonucleic acid (DNA) methylation events usually existed in the CpG dinucleotides and especially in gene promoters. DNA methylation was a major cause of gene silencing and normally appeared in X-chromosome inactivation, imprinted genes, and tumor-suppressor genes. For instance, mutation of CDH1 gene was the main cause of hereditary diffuse GC, and DNA methylation often occurred to inactivate this gene. Therefore, DNA methylation might be considered as a good marker to assess the tumorigenesis, which was also beneficial for cancer diagnosis and treatment.Gastric cancer (GC) is one of the most common malignancies and remains the second most common cause of cancer-related death worldwide.6-methyguanine DNA methyltransferase (MGMT), a DNA repair protein encoded by MGMT gene that located at 10q26, removed cytotoxic and mutagenic adducts from the O6-guanine of DNA. Alkylation at the O6 position of guanine in DNA contributed a lot to the emergence of gene variation in cancers, because thymine tended to replace cytosine to pair with O6-methylguanine in the replication of DNA, which eventually resulted in the G > A mutation. MGMT protein could protect DNA from methylation damages, thus it had the ability of tumor inhibition. In contrast, it might produce chemoresistance to anticancer treatment, therefore leading to treatment failure. The loss of MGMT expression was mainly due to epigenetic event other than the deletions or rearrangements of the gene. The hypermethylation of the CpG islands on MGMT gene promoter significantly silenced the MGMT gene. In recent studies, significant associations between MGMT promoter hypermethylation and glioblastoma, breast cancer, and colorectal cancer were found. Many studies evaluating the correlation of MGMT promoter hypermethylation with GC were also performed. However, limitations of sample type, sample size, and race lowered the reliability of the results. Hence, in order to clarify the association between GC pathogenesis and MGMT promoter hypermethylation, we performed this meta-analysis.O22.1PubMed, Web of Science, Embase, and CNKI databases were searched up to September 2016 using the search terms “Stomach Neoplasms[Mesh],” “gastric cancer,” “gastric tumor,” “gastric carcinoma,” “MGMT protein, human” [supplementary concept], “MGMT,” and “‘Methylation’[Mesh].” The references of included articles and relevant review literatures were searched and scanned to acquire additional eligible studies. In this retrospective meta-analysis, institutional review board approval was not required.2.2Published studies included must meet the following criteria: studies that evaluated the association of MGMT promoter methylation with GC risk or clinical characteristics, studies that provided sufficient data on the promoter methylation of MGMT in control group and GC group, and all cancer patients were diagnosed according to the diagnostic criteria of GC. If studies did not meet these inclusion criteria, they were excluded.2.3Two investigators independently extracted the relevant data from included studies according to the inclusion criteria. Any discrepancy was discussed and settled by consulting with the team. The following information were collected in this meta-analysis: name of the first author, publication year of article, country of subjects, ethnicity of subjects, the number of controls and cases, GC type, clinical information, detection method of methylation, and sample type.2.4http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp) table. The NOS criteria included 3 parts: subjects selection, 0 to 4 scores; comparability of subjects, 0 to 2 scores; and exposure of subjects, 0 to 3 scores. The scores range from 0 to 9, and a score of ≥7 indicated a high-quality study.Two investigators independently assessed methodological quality of included articles using the Newcastle–Ottawa Scale (NOS)  and 95% confidence intervals (95% CIs) were calculated using STATA 12.0 software  software. The degree of heterogeneity present in different studies was performed based on Cochran Q statistic and I2 tests. When a P value of <.05 or I2 > 50% was presented, significant heterogeneity was considered existing among studies, and the random effects model was used to calculate the ORs and 95% CIs; otherwise, the fixed-effects model was applied.,16 Funnel plots were also drawn to investigate the potential publication bias of included studies. If funnel plots were distributed approximately symmetrically, results indicated an absence of obvious publication bias. Begg and Egger tests were also carried out to observe the publication bias. Furthermore, subgroup analysis based on sample subtype and race was performed to lower the between-study heterogeneity. Sensitivity analysis was conducted by excluding individual study to assess stability of results, and outlying studies were removed.In order to evaluate the strength of the association between 33.1The original searching yielded a total of 32 publications using the search terms. Thirty-one publications were obtained after 1 duplicated article was removed. Through reading title and abstract, 24 publications were remained. Finally, 16 articles (17 studies) with 1736 cases and 1291 controls were included in this meta-analysis by scanning the full texts. In these publications, 14 studies had sufficient data of both controls and cases group, and 3 studies only had sufficient data of cases. Thus, the remaining 3 studies were used to observe the association between MGMT promoter hypermethylation and clinical characteristics of GC. Moreover, 8 studies considered adjacent tissue as control group, while 4 studies put normal tissue as control group. All studies on hypermethylation of MGMT promoter were genotyped in GC tissue except 2, which used blood sample. On the other hand, 2 studies were in Asians and 15 studies were in Caucasians . In the subgroup analysis based on sample subtype, the heterogeneity significantly lowered in blood sample and normal tissue . However, significant associations were still found in the subgroup analysis . In the subgroup analysis based on GC subtype, the MGMT promoter hypermethylation significantly increased the risk of gastric adenocarcinoma . The results of subgroup analysis based on race, displaying on the forest plot, indicated that the race was one of the sources of heterogeneity. In addition, the frequency of MGMT promoter hypermethylation in M1-stage of GC was significantly higher than those in M0-stage . In a separate analysis, there was a significant association of MGMT promoter hypermethylation with lymph node (LN) metastasis of gastric tumor , especially in Asians . The probability of occurrence of diffuse GC was 1.81 times than that of intestinal GC . In the analysis for TNM-stage, the MGMT promoter hypermethylation accelerated the progress of GC . Moreover, no significant associations of MGMT promoter hypermethylation with tumor differentiation, T-stage, and gender were detected . The results indicated that sample type and race were not the main source of heterogeneity, especially in Asians. However, the 2 factors, cancer type and ethnicity, still lowered the heterogeneity among studies according to the subgroup analysis.Although a significant association of MGMT promoter hypermethylation with GC risk was found, heterogeneity among studies was found. Considering the existence of heterogeneity, in order to lower the heterogeneity present in different studies, subgroup analysis based on sample subtype and race was performed in this meta-analysis. Furthermore, meta-regression analysis was also carried out to find the source of heterogeneity   Fig. . In the 05) Fig. .4–19 It has been observed that almost half of the tumor suppressor genes were inactivated by gene-promoter hypermethylation in cancers. Therefore, many studies have focused on the discovery of gene-promoter hypermethylation and have found many epigenetic alterations such as p16INK4a, CDKN2B/p151NK2b, and p14ARF. Increasing evidence suggested that different tumors often had different epigenetic alterations such as clustering of gene hypermethylation on the CpG island methylator phenotype. Previous study has found that concordant hypermethylation of multiple locus occurred in 31% GC; however, hypermethylation of independent locus could not satisfactorily predict the risk and prognosis of GC. In fact, no ideal tumor markers for the detection of GC were found in present studies. However, these biomarkers were very important to screen, diagnose, and determine the stage and metastasis of GC. If methylation microarrays could be conveniently used in GC, it might be beneficial for the diagnosis and therapy of GC patients. But before this, a lot of gene methylation alterations should be found and verified in different populations to determine specificity strongly.Traditional genetic studies on GC have found many gene variations such as single-nucleotide polymorphism, copy number variation, and indel.hMLH1 gene was often found in GCs and had a significantly association with the loss of hMLH1 expression. Of note, CDH1 promoter hypermethylation had a high frequency in the diffuse histological type. It has been reported that this hypermethylation alteration contributed a lot to LN metastasis of GC.CDH4 gene hypermethylation has also been found in a higher frequency in GC tissue than adjacent mucosa, and it might have a significant association with progression of GC. Furthermore, APC, H-Cadherin, DAPK, RASSF1A, XIAP, VLDLR, RUNX3, RARβ, CASP8, and MGMT had abnormal epigenetic variations in the development of GC. Although previous studies have found that the hypermethylation of MGMT gene accounted for 31% in GC patients, the association between MGMT promoter hypermethylation and risk of GC was inconsistent. Thus, this meta-analysis was performed by summarizing the existing data to observe the correlation.Currently, many genes hypermethylated in human cell lines and primary tumors have been found. For instance, promoter hypermethylation of ,37,40 while 11 studies observed a significant association of MGMT promoter hypermethylation with risk of GC.–36,38,39 In the calculation of total ORs, the 3 studies contributed to 21.61% of weight; thus, it was necessary to conduct a meta-analysis to determine the association between MGMT promoter hypermethylation and GC risk. From the overall analysis, the MGMT promoter hypermethylation might significantly increase the risk of GC. According to the subgroup of gastric adenocarcinoma, a significant result was also detected, and the carriers of MGMT promoter hypermethylation have 3.47 times higher risk of GC than those of hypomethylation. Furthermore, the frequency of MGMT promoter hypermethylation between males and females did not indicate significant difference . This result showed that the occurrence of MGMT promoter hypermethylation was not related to the gender. Stomach carcinoma might be classified into the diffuse cancer and intestinal cancer. These 2 types of GCs had different pathways and mechanisms. The development of intestinal cancers could be divided into 4 steps: atrophic gastritis, intestinal metaplasia, dysplasia, and GC, while no clear steps of development existed in the diffuse cancer. On the basis of this meta-analysis, different frequencies of MGMT promoter methylation were also found between diffuse and intestinal types. The frequency of MGMT promoter hypermethylation in diffuse type was significantly higher than that in intestinal type. In the included studies on Lauren classification, only 1 study had a same significant result. The significant association of MGMT promoter hypermethylation with distant metastasis and LN metastasis was also observed in the forest plot. However, the 2 significant associations were only detected in Asians, and no between-study heterogeneity was observed. Therefore, the MGMT promoter hypermethylation might encourage the metastasis of gastric tumor in Asians and decrease the survival rate of GC. The result in Caucasians might not be accurate due to the small number of included studies; thus, more large-scale and well designed studies were still needed to be conducted. In the meantime, the GC patients’ number of MGMT promoter hypermethylation in stages III to IV was higher than the number of MGMT promoter hypermethylation in stages I and II. Finally, these results demonstrated that the MGMT promoter hypermethylation had a crucial role in the development of GC. Although heterogeneity was detected in the overall analysis, it was significantly decreased after subgroup analysis and random effects model were conducted. According to the results of meta-regression, ethnicity and cancer subtypes were not the main causes of heterogeneity, and other factors might result in the heterogeneity. Furthermore, no significant publication bias was shown on the basis of Begg and Egger tests. Therefore, although some limitations existed in the meta-analysis, these results might provide a direction that MGMT promoter hypermethylation might contribute a lot to the pathogenesis of GC.From the forest plots, 3 studies did not show a significant correlation, Therefore, to a large extent, the negative result was same with most studies.In addition to these positive results, negative results were also shown in this meta-analysis. No significant associations of MGMT promoter hypermethylation with gastric tumor differentiation were found. In previous studies, the association was contradictory in included studies that were conducted in different sample sizes and races. For tumor differentiation, 1 study indicated that MGMT promoter hypermethylation encouraged the tumor differentiation.I2 = 84.1%, P = .000; for Caucasians, I2 = 66.8%, P = .083; for Asians, I2 = 0.0%, P = .376).In this meta-analysis, we applied the NOS table to evaluate the methodology quality of literatures. The high scores of studies ensured the efficiency and accuracy of statistical analysis. Finally, subgroup based on sample size, sample type, and race was conducted to reduce the heterogeneity, and the heterogeneity lowered obviously. The heterogeneity decreased a lot after subgroup analysis on the basis of race performed, especially in gastric adenocarcinoma (for total, Some limitations still existed in this meta-analysis. First, selection bias was inevitable due to restriction to articles in English or Chinese language. Second, although subgroup analysis and meta-regression were conducted, heterogeneity present in different study still existed. This result might be caused by individual difference; GC subtypes; or other environmental factors such as diet, smoking, and work environment. Third, the small sample size of studies involving clinical information restricted the statistical power. Fourth, the studied population in these articles only included Caucasians and Asians, and the Asians accounted for a lot. Hence, these results might not represent the overall tendency in the total crowd. Finally, the accuracy of results might be influenced by the small number of subjects and different sample type. Although the number of subjects was moderate, but it was still small after the subgroup analysis based on race and sample type was performed.MGMT gene-promoter hypermethylation was significantly associated with the risk of GC, especially in Asians and gastric adenocarcinoma. Furthermore, MGMT gene-promoter hypermethylation might promote the distant metastasis and LN metastasis of gastric tumor.In conclusion, the results of this study indicated that


In pesticide exposure groups such as farmers, the risk of prostate cancer was increased, although the report of the cause of evidence is limited. We selected chlorophenol compounds as an important group of the contaminated pesticide with highly toxic 2, 3, 7, 8-tetrachlorodibenzo-p-dioxin (TCDD). This meta-analysis, the risk of death from prostate cancer was analyzed.PubMed, Scopus, Scholar Google and web of Sciences until 2016 were searched. The standardized mortality rate (SMR) and 95% confidence intervals (CI) were obtained from the studies. We tested statistical heterogeneity with Cochrane Q test and I2 index. Egger test was used for evaluating publication bias. Random or fixed-effects models and meta-regression were also used in our analysis. Moreover, Cochrane tool was used to assess the risk of bias.P=0.027). Some biases are more probable in these studies such as confounding by indication, loss to follow up and misclassification.Five available papers consist of 28706 exposed populations were assessed. Overall standardized mortality rate as combined result of prostate cancer risk from the fixed model was 1.2  1.02 to 1.42, A contaminated pesticide with dioxins between other pesticides is an important risk factor for prostate cancers. Prostate cancer is the most common cancer in men the worldwide , 2. Age Much of the research demonstrate mortality of soft-tissue sarcoma and non-Hodgkin's lymphoma in chlorophenol exposure contaminated to TCDD , 16, wheIn this study, we conducted a meta-analysis of the literature focused on the association between dioxin exposure among workers in manufacture of pesticides producing and prostate cancers.We collected all studies of prostate cancer in chlorophenol compounds production through 2016 by using PubMed, Scopus, Scholar Google (until first 20 pages), and Web of Sciences databases using different language. The key terms for searching were “pesticide”, “chlorophenol”, “phenoxy acetic acid”, “prostate”, “cancer”, “neoplasm” and “dioxin” with different combinations. The references of articles were manually searched to find related publications.Publications that report the risk of prostate cancerPublications including prostate cancer risk in workers of production factory with chronic exposure to CPs.Prostate cancer risk reporting in accidental exposure was excluded.Overall, 784 studies were included. After checking titles and abstracts of all publications 25 studies were selected which reported cancer in CPs production factory. After reviewing the full text of these papers, 13 studies were selected based on the following inclusion criteria:By comparing the full text of these 13 studies, six publications –26 were Therefore, five published articles –29 were Response rate and 95% confidence interval, the country where study was accomplished, period of study, year of publication, study design , sample of exposed group, co-morbidities, time of exposure, number of authors and journal impact factor (IF) in the year of publication was extracted from each study. In this study, response rate of exposed group was classified by latency of effect and type of component exposure. We asked authors some other information, which there was not in the papers. They E-mailed some of our questions except for place of the study .www.strobe-statement.org available at 8 April 2013]. In order to assess the risk of bias , Cochrane tool was used. We evaluated selection bias , information bias , confounding bias  and loss of follow up for each study. Overall assessment risk of bias for each study classified as green (low risk of bias), yellow (unclear risk of bias) or red (high risk of bias) according to each key criteria.The quality of study was assessed by STROBE guideline [2 factor where I2 < 25% have low, between 25% to 50% moderate, and I2 >50% high degree of heterogeneity  a fixed-effects model of Mantel-Haenszel test was used to calculate the meta SMRs. Finally, the random effects model based on DerSimonian and Laird method was evaluated when studies were heterogeneous.Standardized mortality ratios and 95% CI were evaluated for each study. To test comparability among studies; we tested the homogeneity between the studies with Cochrane Q test. The degree of heterogeneity was estimated using Iogeneity . If heteValidity of meta-analysis was assessed by publication bias based on Egger test and Begg’S funnel plot. Meta-regression used for exploring the significant sources of heterogeneity while we considered all probable variables that could change tau index. Begg’s funnel plot was characterized by plotting the logarithm of prostate cancer SMR versus the standard error where an asymmetry around Begg’s funnel plot specifies sources of heterogeneity.Five papers with six cohort reports published data on the association of dioxin exposure in CPs production and prostate cancer that total of 28706 exposed populations involved in the analysis. Fifty-two percent of populations included in the studies were from the USA and 48% from Europe. The study , 20 apprA publication of IARC study covering 36 previous studies from 12 countries when it followed prostate cancer in pesticide production from1939 to 1992 . Loss toP-value=0.93). The degree of heterogeneity was estimated low based on the I2 <25%. The fixed effect model was applied in this analysis and produced a meta-rate ratio of 1.2  where yielded 20% additional risk for death from prostate cancer in exposed group (No Significant heterogeneity existed among the combination of six SMRs reported in the studies (ed group .P=0.03). Meta-analysis excluding the study .Once we excluded the study  which fohe study  revealedhe study , 20, thrhe study , 28, 29 he study , 28. CPsWe did not demonstrate any significant relationship between each one of factors like period of study, year of publication, study design, sample of exposed group, time of exposure, number of authors, journal impact factor (IF) and latency of exposure which sample had exposure with them and SMR of prostate cancer. However, when we sorted studies based on their midpoint year of running the study, there was an increasing trend for SMR of prostate cancer, except the newest study , which hIn recent years, the incidence of the risk of prostate cancer mentioned as one of the most frequently reporting , 2. OccuBy bias assessment, only one report can be reliable among included studies in meta-analysis. Different types of selection bias were less probable. Therefore, the results of these studies are much probably generalizable to other cases with prostate cancer at risk of exposure to dioxin compounds. However, internal validity of these studies due to different types of information bias is under question. Sensitivity analysis in these studies could be a solution for compensating for loss to follow up.In the previous meta-analysis, the association between pesticide exposure in different occupation  and prosOur analysis supports a significant of 20% increase in SMR of prostate cancer for population with dioxin exposure. Early historical cohort studies by the International Agency for Research on Cancer were covering 36 previous studies in the USA and Europe from 1939 to 1992 . Rate raIn different exposure of chlorophenols and phenoxy acetic acids, CPs exposures significantly highlighted development of prostate cancer when a non-significant increase of prostate cancer was reported in workers exposed to CPs pesticide , 20.One reason for absence of significant relationship between each one of factors like period of study, year of publication, study design, sample of exposed group, time of exposure, number of authors, journal impact factor (IF) and latency of exposure and SMR of prostate cancer could be related to low number of investigated studies in our meta-analysis.However, additional unknown factors might also be applied since the lack of smoking and dietary habits, and intake of alcohol information on the incidence of prostate cancer may play a role and we had no information about them in these studies as a limitation. Extensive in vitro and in vivo toxicity testing of dioxin-contaminated chlorophenols has shown activity of androgen receptor modulators effect for these compounds which stimulates cell proliferation , 34. FinOur findings has strengthened the evidence of prostate cancer in occupational exposure to chlorophenols. These results confirmed increase of death risk from prostate cancer.Ethical issues  have been completely observed by the authors.


A custom-by-design 48-Plex SNPscan Kit was used to determine the genotypes of MTHFR rs3753584 T>C, rs9651118 T>C, rs1801133 G>A, rs4846048 A>G and rs4845882 G>A polymorphisms in 387 CRC patients and 1,536 non-cancer controls. The results revealed that MTHFR rs1801133 G>A polymorphism was associated with a decreased risk of overall CRC. While MTHFR rs4845882 G>A polymorphism conferred an increased risk to overall CRC. In a stratified analysis by CRC region, we found MTHFR rs3753584 T>C and rs9651118 T>C polymorphisms were associated with the increased risk of colon cancer. In addition, a significantly increased risk of rectum cancer associated with MTHFR rs3753584 T>C polymorphism was overt. However, MTHFR rs1801133 G>A polymorphism conferred a decreased risk to colon cancer. In conclusion, findings of the present study reveal that the tagging polymorphisms in MTHFR gene  are associated with the increased risk of CRC. However, MTHFR rs1801133 G>A polymorphism confers a decreased risk to CRC. Additional studies with larger sample size are needed to confirm these findings.The aim of this case-control study was to assess the relationship between the tagging polymorphisms in  Colorectal cancer (CRC) is the fourth most frequent type of malignancy among females and the fifth most frequent type among males in China, accounting for 160,600 and 215,700 cases in 2015, respectively –9. AccumSeveral studies have found association between circulating or dietary folate level and the risk of CRC –12. A loMTHFR gene locates on 1p36.22. The gene encodes a 74.6-kDa protein containing 656 amino acids. MTHFR is also linked to purine synthesis; therefore, plays a vital role in DNA synthesis by the provision of essential nucleotides  were analyzed and selected from the data of Chinese Han individuals in Beijing (CHB) via the HapMap Project (http://hapmap.ncbi.nlm.nih.gov/index.html.en) [MTHFR tagging SNPs is presented in Table The tagging SNPs of html.en) . The dethtml.en) . The infEthylenediamine tetraacetic acid-anticoagulated intravenous blood was donated by every participant. Using the Promega DNA Blood Mini Kit , genomic DNA was extracted from peripheral lymphocytes by the standard experimental protocol.MTHFR rs3753584 T>C, rs9651118 T>C, rs1801133 G>A, rs4846048 A>G and rs4845882 G>A polymorphisms as described in previous studies [A custom-by-design 48-Plex SNPscan Kit , double ligation and multiplex fluorescence PCR , was per studies , 24. A tP < 0.05 (two–tailed) was considered to be a statistical significance. The quantitative variables were expressed as means ± standard deviation (SD). Student's t-test was harnessed to evaluate the difference of age between CRC patients and controls. Additionally, we used 2χ test to examine the differences in age, sex, smoking status, alcohol consumption and the frequencies of genotype between patients with CRC and controls. The Hardy-Weinberg equilibrium (HWE) equation was used to assess whether the proportion of MTHFR tagging SNPs genotypes obtained was in accordance with the expected value. An online calculator (http://ihg.gsf.de/cgi-bin/hw/hwa1.pl) was harnessed to calculate the P value of HWE [MTHFR rs3753584 T>C, rs9651118 T>C, rs1801133 G>A, rs4846048 A>G and rs4845882 G>A genotypes with CRC risk was estimated by crude/adjusted odds ratios (ORs) and 95% confidence intervals (CIs). The power of the present study (α= 0.05) was evaluated by the Power and Sample Size Calculator (http://biostat.mc.vanderbilt.edu/twiki/bin/view/Main/PowerSampleSize).We used SAS statistical software, version 9.4  for data analysis and a e of HWE . The rel