RCTs) and observational studies that evaluated the association between pioglitazone and bladder cancer risk. Conventional and cumulative meta‐analyses were used to calculate the odds ratio (OR) with 95% confidence interval (CI). A restricted spline regression analysis was used to examine the dose–response relationship with a generalized least‐squares trend test. We included two RCTs involving 9114 patients and 20 observational studies . An increased risk of bladder cancer in patients treated with pioglitazone versus placebo was noted from RCTs . In observational studies, the increased risk of bladder cancer was slight but significant among ever‐users of pioglitazone versus never‐users , which appeared to be both time‐ (P = 0.003) and dose‐dependent (P = 0.05). In addition, we observed the association differed by region of studies  or source of funding (sponsored by industry or not). Current evidence suggests that pioglitazone may increase the risk of bladder cancer, possibly in a dose‐ and time‐dependent manner. Patients with long‐term and high‐dose exposure to pioglitazone should be monitored regularly for signs of bladder cancer.Current evidence about the association between pioglitazone and bladder cancer risk remains conflict. We aimed to assess the risk of bladder cancer associated with the use of pioglitazone and identify modifiers that affect the results. We systematically searched PubMed, Embase, and Cochrane Central Register of Controlled Trials from inception to 25 August 2016 for randomized controlled trials ( In 2015, one ten‐year interim analysis in a large observational study of the Kaiser Permanente Northern California (KPNC) database, which was required by US FDA, found that use of pioglitazone was not significantly associated with increased risk of bladder cancer in the United States Because of the ongoing safety concerns, there is a critical need for a rigorous assessment of the bladder safety of pioglitazone. Furthermore, several effect modifiers  may contribute to the risk of bladder cancer; however, they have not been fully explored in previous meta‐analyses Following the PRISMA guidelines We included randomized controlled trials (RCTs) fulfilling the following inclusion criteria: (1) parallel‐group design; (2) pioglitazone as the intervention versus a control treatment including nonthiazolidinedione therapy, nonpioglitazone therapy, or placebo; and (3) reporting the events of bladder cancer. In addition, we also considered observational studies (cohort or case‐control studies) reporting on the risk of bladder cancer associated with use of pioglitazone versus never use of pioglitazone/thiazolidinedione. We excluded the conference abstracts as well as the articles without reporting the bladder cancer outcome.Two reviewers (HT and WS) independently checked titles and abstracts to exclude articles that clearly did not fulfill the inclusion criteria. Any potentially eligible studies were further assessed by retrieving full texts. A standardized data extraction form was used to collect the following information: study design, drug use, region of study, characteristics of participants, selection criteria, exposure definition, controlled covariates, funding, and data on bladder cancer risk , adjusted risk ratio (RR), and adjusted odds ratio (OR)).Two reviewers (HT and WS) independently used the Cochrane risk‐of‐bias tool and the Newcastle‐Ottawa quality assessment scale (NOS) to assess the quality of RCTs and observational studies, respectively I2 statistic, with I2 of <25%, ≥25% and <75%, and ≥75% indicating low, medium, and high heterogeneity, respectively We pooled the data for RCTs and observational studies separately. For RCTs, a Peto's OR with 95% confidence interval (CI) was used to pool the outcome data due to rare events of bladder cancer We tested possible dose and time responses of risk of bladder cancer using restricted cubic splines random‐effects meta‐analysis, with three knots at the 10th, 50th, and 90th percentiles of the distribution. We also used the method described by Greenland and Longnecker P < 0.05 was considered statistically significant.All statistical analyses were performed with STATA . A two‐tailed Of 5224 citations identified from electronic databases, 128 studies were potentially eligible after initial title and abstract evaluations. After further assessment through retrieving full texts, 22 studies proved eligible, including two RCTs and 20 observational studies (12 cohort studies and eight case–control studies). No additional studies were identified in the manual search. The process of study selection with reasons for exclusion is presented in Figure .Two large randomized, placebo‐controlled trials reporting the outcomes of bladder cancer were eligible and included I2 = 0%; Fig. These two trials reported a total of 40 cases of bladder cancer among 9114 patients (raw event rate 0.4%), with 26 cases among 4544 patients (raw event rate 0.6%) treated with pioglitazone and 14 cases among 4570 patients (raw event rate 0.3%) treated with placebo. Meta‐analysis of these two trials showed a borderline increase in the risk of bladder cancer when pioglitazone was compared with placebo , with no evidence of between‐study heterogeneity (Twenty studies (12 cohort studies and eight case–control studies) involving 4,846,088 individuals reported adjusted estimates of bladder cancer I2 = 31.3%; Fig. P = 0.09), Begg's test (P = 0.97), or visual inspection of the funnel plot , with moderate between‐study heterogeneity  and dose–response relationship (P = 0.04).The results from stratified analyses are presented in Table P = 0.77). Similarly, there was little evidence that the risks differed depending on whether pioglitazone compared with never use of any thiazolidinedione , never use of pioglitazone , or other comparison groups. An increased risk of bladder cancer was observed in the studies with smoking adjusted , studies without receiving the funding from industry , and cohort studies . There was no significant difference between men and women (P = 0.47).The stratified analysis showed that the increased risk of bladder cancer varied depending on the populations from Europe , Asia , or United States , but not significant difference among these three groups  from PROactive were diagnosed within the first year of treatment Interestingly, our results indicate that the increased risk of bladder cancer associated with pioglitazone was detected in European populations, but not in US or Asian populations. Differences among ethnicities might be involved. A higher rate of bladder cancer was observed in Caucasians when compared with Blacks and Asians Six meta‐analyses were conducted to evaluate the association between pioglitazone and risk of bladder cancer Table  13, 14, Our study has several strengths. First, we systematically searched electronic databases and included both randomized trials and observational studies to evaluate the association between pioglitazone and risk of bladder cancer. Second, we explored sources of heterogeneity by performing meta‐regression meta‐analysis and several prespecified stratified analyses. Third, we assessed the robustness of evidence by carrying out cumulative and sensitivity analyses. Finally, a restricted spline regression analysis was used to examine the dose‐response relationship with a generalized least‐squares trend test. We also acknowledge that our meta‐analysis has several limitations. First, our results should be interpreted with caution due to between‐study heterogeneity in exposure and population. The duration of follow‐up, duration of diabetes, and other drugs used varied across studies, although several prespecified stratified analyses were performed to minimize the heterogeneity. The cumulative dose and cumulative duration of pioglitazone use were different. Furthermore, not every study included these results for inclusion in our meta‐analysis. Second, different observational studies adjusted for different sets of confounders, although we used the risk estimates that fully adjusted for the covariates and performed a stratified analysis by major factor adjusted (smoking status). Third, several studies performed a nested case–control in a cohort and reported both results In summary, our meta‐analysis of currently available evidence suggests that pioglitazone is associated with slightly but significantly increased risk of bladder cancer, indicating a time‐ and dose‐response relationship. The risk of bladder cancer associated with pioglitazone is various and depended on ethnicity and choice of comparison group. Our updated and detailed information suggests cautions and closely monitoring signs for bladder cancer in patients with T2DM, who use pioglitazone, especially those requiring higher doses and longer duration of treatment. However, further well‐designed and well‐conducted studies that clearly define study population and exposure, and adequately adjust confounders are needed to confirm our findings.The authors declare that they have no conflict of interest.Table S1. Characteristics of included RCTs.Table S2. Characteristics of included observational studies.Table S3. Quality assessment of observational studies.Table S4. Summary results of previous meta‐analyses.Figure S1. Risk of bias assessments of included RCTs.Figure S2. Sensitivity analysis of the association between pioglitazone use and risk of bladder cancer when omitting each study successively based on adjusted data from observational studies.Figure S3. Sensitivity analysis of the association between pioglitazone use and risk of bladder cancer by including the most recent study only based on adjusted data from observational studies.Figure S4. Funnel plot of the association between pioglitazone use and risk of bladder cancer based adjusted data from observational studies.Click here for additional data file.


Previous epidemiological studies have provided inconsistent conclusions on the effect of coffee consumption in the development of myocardial infarction (MI). The aim of the study was to evaluate the influence of coffee consumption and its potential dose-response patterns on the risk of developing MI.Three databases were searched for evidence of eligible studies. A random-effects model was used to pool the fully adjusted odds ratios (ORs) and the corresponding 95% confidence intervals (CIs). Dose-response analysis was performed to show the effect of each cup increased in daily coffee drinking on the risk of MI.Seventeen studies involving 233,617 participants were included in our study. The association between coffee consumption and risk of MI did not show statistical significance when pooling the outcome data for the coffee consumption categories of 1~2 vs. < 1 cup per day  and 2~3 vs. < 1 cup per day . Compared with < 1 cup, daily drinking of 3~4 cups and > 4 cups of coffee were significantly associated with the risk of MI, and the pooled ORs (95% CIs) were 1.40 (1.11–1.77) and 1.48 (1.22–1.79), respectively. The dose–response analysis showed a “J–shaped” curve relationship of the risk of MI with coffee consumption.Daily drinking of more than three cups of coffee was associated with a significantly increased risk of MI. This positive association was only found in men but not in women. The impact of gender on this association should be further evaluated. Coffee is one of the world's most popular beverages, and it has been implicated in the development of cardiovascular diseases (CVDs) , 2. As tAlthough previous meta-analyses have concluded the association between coffee drinking and incidence of coronary heart disease  and overThe selection process of eligible studies is shown in detail in a flow diagram Figure . In the The characteristics of each included article were shown in 2 = 17.6, p = 0.243). In the subgroup analyses, the non-significant association was not changed by different sexes, coffee types, study locations, study designs, and NOS scores  Figure . The pooes Table .2 = 27.4, p = 0.162). There was still no significant association in the subgroup analyses of sex, coffee type, study location, study design, and NOS score  Figure . The hetre Table .2 = 57.5, p = 0.005). In the subgroup analyses, the positive association between coffee consumption and risk of MI was only found in the pooling result for men , but not for women  or both sexes  (Table p = 0.02). Moreover, this significant association was only found in studies that used all types of coffee consumption as the exposure of interest but not in studies that used caffeinated coffee or decaffeinated coffee, and in studies with NOS score of 7 but not in studies with NOS score of 9 and 8 (Table p = 0.28) and design (p = 0.09) were non-significant. The sensitive analysis did not significantly change the results by omitting one study in turn, and pooled results of OR range from 1.31 (95% CI: 1.05–1.63) to 1.51 (95% CI: 1.24–1.84)  Figure , and the24–1.84) .2 = 68.5, p < 0.001). Similarly, the significantly association was not observed in studies that included only women in the subgroup analysis  (Table p = 0.04). In addition, the higher risk was not significant in the studies that used caffeinated coffee as the exposure of interest, in studies that used a cohort design, or in studies with NOS score of 9 (Table p = 0.33), design (p = 0.26), and study quality (p = 0.41) were non-significant. The sensitive analysis did not significantly change the results by omitting one study in turn, and pooled results of OR range from 1.41 (95% CI: 1.18–1.70) to 1.55 (95% CI: 1.29–1.86)  Figure , with ev29–1.86) .p = 0.034). The best fitting model showed a “J-shaped” curve  and Begg-Mazumdar test  also suggested that no publication bias was observed in the meta-analyses.The studies were distributed fairly symmetrically about the combined effect size in funnel plots Figure , which iOur study included 17 articles involving 233,617 participants and 10,510 patients with MI. Compared with less than 1 cup, coffee consumption of 1~2 cups and 2~3 cups per day did not significantly increased with the risk of MI; however, coffee consumption of 3~4 cups and > 4 cups per day was positively related with the risk of MI. This means that low to moderate coffee intake was not associated with an increased risk of MI, but heavy or very heavy intake significantly increase the risk. A dose-response relationship was observed between coffee consumption and the risk of MI, with a “J-shaped” relationship. Furthermore, the positive association was not changed in the sensitivity analyses and there was no publication bias in all the pooled results. Current dietary guidelines recommend moderate coffee intake , which wPrevious meta-analyses have assessed the association between coffee drinking and coronary heart disease risk , 33, 34.The most of individual studies only included male participants or female participants, and although some studies included both men and women, few studies have conducted stratified analysis by sex. In our meta-analysis, we conducted a subgroup analysis by sex and assessed the impact of sex on the pooled results. In the subgroup analyses by sex, we found that 1~2 cups and 2~3 cups per day of coffee consumption did not increase the risk of MI in both men and women, comparing with less than 1 cup per day; however, the point estimates were over 1 for men  and below 1 for women . In addition, 3~4 cups and > 4 cups per day of coffee consumption significantly increased the risk of MI in men  but not in women , comparing with less than 1 cup per day. Moreover, the interaction for sex and the risk of MI was significant. These results indicated that MI induced by coffee consumption is more serious for men than for women. This may be due to the fact that the number of estimates for women is less than men, which leads to a low statistical power. Other studies also reported different results on coffee consumption and CVD risk by different genders , 36, 37.1 and A2A adenosine receptors ”, “Myocardial Infarction [MeSH]”, “coffee”, “coffee intake”, “coffee consumption”, and “myocardial infarction”. The search strategy was limited by study design, and only cohort studies and case-control studies were included. The references of all retrieved articles and recent review were also manually screened. No attempt was made to find articles in languages other than English. We contacted the authors for additional data when necessary.A study was eligible for inclusion if the following criteria were met: (1) a cohort or case-control design; (2) examination of coffee consumption as the exposure of interest; (3) determination of MI as the outcome of interest; and (4) reporting the relative risks with their 95% confidence intervals (CIs) of MI relating to three or more categories of coffee consumption. The studies about review research, animal experiment, and mechanistic research were excluded.Two investigators extracted the data independently using a predefined data collection form, with disagreements being resolved by consensus. Relevant data included the first author's surname, publication year, location of the study, study design, gender of the participants, age of the participants, sample size of the study, number of MI events, ascertainment method of coffee consumption, ascertainment method of MI, coffee type, coffee consumption categories, most fully adjusted relative risks and corresponding 95% CIs for every category of coffee consumption, and covariates adjusted in the multi-variable analysis.The quality of each included study was assessed by two investigators, using the Newcastle-Ottawa Scale (NOS) recommended by Wells and colleagues . The NOS2 statistic was used to examine the heterogeneity between the studies (I2 > 50% was indicated as significant heterogeneity) [Data was classified into five categories of coffee consumption as follows: < 1 cup per day, 1~2 cups per day, 2~3 cups per day, 3~4 cup per day and > 4 cups per day. An approach of generic inverse-variance was used to pool the outcome data for the coffee consumption categories of 1~2 vs. < 1 cup per day, 2~3 vs. < 1 cup per day, 3~4 vs. < 1 cup per day and > 4 vs. < 1 cup per day with a random-effects model. Igeneity) . Subgrougeneity)  and Begggeneity)  were useIn the dose-response analysis, a generalized least squares trend estimation model was used to compute the specific slopes , 61. ForP values less than 0.05 were considered as significant.Statistical analyses were conducted using the Stata (version 12.0). All


Here, our aim is to examine the possible effect of folate, homocysteine, and vitamin B12 levels on the risk of colorectal polyps by means of meta-analysis based quantitative synthesis. According to our inclusion/exclusion criteria, a total of 13 case–control studies were enrolled. The P-value of the association test, standard mean difference (SMD), and 95% confidence interval (CI) were calculated. Pooled analysis data showed a negative correlation between the risk of colorectal polyps and the levels of serum folate, red blood cell (RBC) folate, or vitamin B12 . Nevertheless, for homocysteine level, we also observed a statistically significant difference between cases and controls in the overall and subgroup analysis of hospital-based control (HB), population-based control (PB), Chinese, Caucasian, or Asian . We found that increased levels of homocysteine may be statistically and significantly related to the risk of colorectal polyps.Several studies have reported an association between levels of folate, homocysteine, and vitamin B Colorectal polyps are abnormal growths and protrusions on the colorectal surface ,2. The m12, also termed as cobalamin, has been found to be involved in homocysteine metabolic reactions, DNA synthesis, mitochondrial metabolism, and semen quality ) OR polyp) OR polyposis) OR colorectal polyps) OR colorectal polyp) OR colorectal polyposis) OR adenomatous polyposis) OR hyperplastic polyps) OR intestinal polyps) OR colonic polyps) OR polyposis coli) AND (((((((((folate) OR serum folate) OR RBC folate) OR red blood cell folate) OR folate status) OR folate metabolism)) OR (((((((homocysteine [MeSH terms]) OR 2-amino-4-mercaptobutyric acid) OR 2 amino 4 mercaptobutyric acid) OR homocysteine, L-isomer) OR homocysteine, L Isomer) OR L-Isomer homocysteine) OR plasma homocysteine)) OR ((((((((((vitamin B 12[MeSH terms]) OR B 12, vitamin) OR vitamin B12) OR B12, vitamin) OR cyanocobalamin) OR cobalamins) OR cobalamin) OR Eritron) OR VB12) OR serum vitamin B12)).12 levels; data of intake of folate or vitamin B12; lack of control data or S.D. The process of database searching and study selection was performed in accordance with the recommendations regarding preferred reporting items for systematic reviews and meta-analyses (PRISMA) [We then independently reviewed and screened the eligible case–control studies using our selection criteria, which were duplicate data; reviews; cases or trials; cell or animal data; meeting abstract or poster; meta-analysis; study of polyps other than colorectal, no mention of folate, homocysteine, or vitamin B(PRISMA) .Next, we performed the data extraction based on the included original case–control studies. A specifically designed table was used to show the detailed characteristics of all studies, including first author name, year of publication, group, number of case–control studies, mean value, S.D. value, race, country, disease type, control source. Missing information is designated ‘NA’ (not available).P-value of association test, pooled SMD, and 95% confidence interval (CI) were synthesized in the overall and subgroup meta-analysis by such factors as race, country, and control source. A two-tailed P<0.05 was considered indicative of statistically significant difference. Additionally, based on Q-statistic and 2I test, the evaluation of interstudy heterogeneity was carried out. P-values of Q-statistic < 0.05 or 2I values > 50% were considered indicative of high heterogeneity. We here used an inverse variance (IV)-weighted, random-effects model.Standard mean difference (SMD) was used as the evaluation criterion of the continuous data with varied measurement units, as in similar meta-analyses . Base onP-value of Begg’s test and Egger’s test < 0.05, and asymmetric funnel plot indicate significant publication bias.Both Begg’s test and Egger’s test were performed to quantitatively judge the possible publication bias. We performed a sensitivity analysis to evaluate the sources of heterogeneity and stability of the data. We removed each case–control study from the analysis one by one and analyzed differences in pooled data. All these tests were performed using STATA software .n=258), reviews (n=256), cases or trials (n=40), cell or animal data (n=39), meeting abstract or poster (n=64), meta-analysis (n=8), polyps that were not colorectal (n=76), or the absence of folate, homocysteine, and vitamin B12 levels (n=56). We then obtained a total of 55 publications with full text for eligibility and removed 31 publications containing the data of intake of folate or vitamin B12, and 11 publications that lacked control data or S.D. data. Finally, a total of 13 case–control studies [We gathered a total of 852 relevant publications from a search of three databases, specifically 201 publications in PubMed, 228 in WOS, and 423 in Embase. Of 852 publications, the following records were excluded: duplicate publications . We also conducted subgroup analyses by country, race, and control source. Similar negative results were obtained (P>0.05). It should be noted that we only show the results of subgroup analysis with more than or equal to three case-control studies in the present study. Forest plots of each subgroup analysis by race are given in To evaluate the association between level of serum/RBC folate and risk of colorectal polyps, six case–control studies with 241 cases and 623 controls were enrolled for meta-analysis of serum folate, while 9 case–control studies with 790 cases and 1268 controls were for RBC folate . CompareP<0.001, SMD = 0.52, 95% CIs = 0.25–0.78). Data from the subgroup analysis of hospital-based control (HB), PB, China, Caucasian, and Asian showed similar positive results . However, no significant difference between cases and controls was observed in the overall or subgroup meta-analyses of vitamin B12 (P>0.05). Thirteen case–control studies involving 879 cases and 1818 controls were enrolled in this meta-analysis regarding the relationship between the level of homocysteine and colorectal polyp risk. Data from the overall meta-analysis  indicate2I > 50.0%, P-value of heterogeneity <0.05), and IV-weighted random effect models were thus used in Cohen statistics. For publication bias, as shown in P=0.024), P-value of Begg’s test and Egger’s test was larger than 0.05 in others, indicating the absence of large publication bias. Begg’s funnel plot for the association of homocysteine level and colorectal polyp risks is shown in 12). In addition, similar pooled ORs were observed in our sensitivity analysis, as shown in 12).Obvious heterogeneity was detected in all the comparisons given above  A higher plasma homocysteine concentration was reported to be associated with the risk of advanced adenoma in female Korean participants, but not in the overall population [Several limitations should be fully considered. (i) As in other meta-analyses, there exists the problem of limited case–control studies for quantitative synthesis. Only one study provided data covering the association between serum vitamin Bpulation , which cpulation . The infpulation . For thi12 is also essential to the conversion of homocysteine into methionine [MTHFR gene for C677T polymorphism have been reported to render enzyme activity lower than in the wild-type genotype [Data regarding several studies support the association between the high homocysteine concentrations in the risk of developing CRC ,43. Howethionine . Homozyggenotype ,17. We s12 levels function in the susceptibility to colorectal polyps.Generally speaking, quantitative synthesis data provide evidence regarding the role of an elevated homocysteine level in the developing colorectal polyp risk. Larger sample sizes were still required to investigate whether serum folate, RBC folate, and vitamin B


P<0.05). Association for SNP rs6983267 in the CCAT2 gene only was noteworthy after multiple test correction (P = 0.001). Site-specific analysis showed association between SNPs rs1799853 and rs2302615 with reduced colon cancer risk only , however neither reached significance threshold following multiple test correction. Meta-analysis of SNPs rs2070959 and rs1105879 in UGT1A6 gene showed interaction between aspirin use and CRC risk ; stratification by aspirin use showed an association for decreased CRC risk for aspirin users having a wild-type genotype  compared to variant allele cariers. The direction of the interaction however is in contrast to that published in studies on colorectal adenomas. Both SNPs showed potential site-specific interaction with aspirin use and colon cancer risk only , with the direction of association similar to that observed for CRC. Additionally, they showed interaction between any non-steroidal anti-inflammatory drugs (including aspirin) use and CRC risk (Pinteraction = 0.01 for both). All gene x environment (GxE) interactions however were not significant after multiple test correction. Candidate gene investigation indicated no evidence of GxE interaction between genetic variants in genes involved in aspirin pathways, regular aspirin use and colorectal cancer risk.Regular aspirin use is associated with reduced risk of colorectal cancer (CRC). Variation in aspirin’s chemoprevention efficacy has been attributed to the presence of single nucleotide polymorphisms (SNPs). We conducted a meta-analysis using two large population-based case-control datasets, the UK-Leeds Colorectal Cancer Study Group and the NIH-Colon Cancer Family Registry, having a combined total of 3325 cases and 2262 controls. The aim was to assess 42 candidate SNPs in 15 genes whose association with colorectal cancer risk was putatively modified by aspirin use, in the literature. Log odds ratios (ORs) and standard errors were estimated for each dataset separately using logistic regression adjusting for age, sex and study site, and dataset-specific results were combined using random effects meta-analysis. Meta-analysis showed association between SNPs rs6983267, rs11694911 and rs2302615 with CRC risk reduction (All  Observational studies have consistently demonstrated an association between regular aspirin or other non-steroidal anti-inflammatory drug (NSAID) use and reduced colorectal cancer (CRC) risk . For CRCet al 2015, investigated potential GxE interactions based on regular use of any NSAIDs (including aspirin) as exposure [a priori biologically-motivated hypothesis to assess association between variants in putative genes and phenotype of interest. Hence, the techniques are complementary approaches to identifying potential sources of GxE interactions within observational studies.Inter-individual variation in the chemopreventive effect on colorectal neoplasia has been in part attributed to germline variations, particularly single nucleotide polymorphisms (SNPs). Whilst a large number of SNPs in aspirin’s pharmacokinetic and pharmacodynamic pathways have been suggested to modulate chemopreventive effect in observational studies , 11–17, exposure . Whilst,To our knowledge, no meta-analysis has been reported in the literature which tested for GxE interaction between SNPs in aspirin pathway genes, regular aspirin use and CRC risk. Therefore, the current study aimed to assess associations of candidate SNPs that putatively modify aspirin’s pharmacological effects by carrying out random effects meta-analysis of GxE interaction between aspirin only use and CRC risk using two large population based case-control datasets: UK-Colorectal Cancer Study Group (UK-CCSG) and NIH- Colon Cancer Family Registry (NIH-CCFR).Individual level data from the UK-CCSG (recruitment 1997–2013) and NIH-CCFR Phase I  case-control datasets were pooled. Both studies enrolled incident CRC cases and healthy population controls , 19.From the UK-CCSG dataset, CRC cases and controls from all three study sites  were included in the analysis. Briefly, cases between the age of 45 and 80 years with histologically confirmed incident CRC and diagnosed in the period of 1997–2000, were identified at all three sites while for 2000–2013 only from Leeds. Patients who had a primary cancer previously, history of coeliac disease, familial adenomatous polyposis, diverticular disease 2 years before current cancer diagnosis, non-adenocarcinoma colorectal cancer or ulcerative colitis diagnosed in previous 3 years were not recruited in the study. Healthy population based controls were identified through patient’s GP practice list. An age and sex matched control with no history of previous cancer at the time of recruitment was identified for each case between 1997 and 2000 at all 3 sites, and following 2000, friends or spouse of cases from Leeds site with no history of cancer at the time of recruitment were collected for the study to complement GP recruitment.From the NIH-CCFR dataset, CRC cases and controls from three study sites  were included in the analysis. Despite availability of epidemiological data on cases and controls from the three other NIH-CCFR sites , these sites were excluded from the analysis due to the absence of genome-wide SNP genotype data for controls. For the current study, incident case proband identified through population based cancer registries recruited between 1997 and 2002 were included in the analysis. Healthy population based and spouse controls were identified through medicare and driver’s license files, telephone subscriber lists and electoral rolls and were randomly selected between 1997 and 2002.http://coloncfr.org/questionnaires.For the UK-CCSG dataset, a research nurse carried out interviews of the study participants either in hospital or at home. All participants completed detailed diet and lifestyle questionnaire called the Food Frequency and Epidemiology Questionnaire , which wAll participants provided written informed consent at the time of the interview and the study design was approved by the Institutional Review Boards at each NIH-CCFR site. The UK-CCSG study as well as the joint analysis of the UK-CCSG and NIH-CCFR data were approved by the Leeds East Research Ethics Committee.Because of sample numbers, only self-reported non-Hispanic white individuals were included in the analysis. A baseline epidemiologic questionnaire containing details about medical history and medication use  was completed by participants in both studies; the measurement instrument differed between the studies overall but were equivalent for assessing NSAID usage .Literature review was carried out using PubMed and Google Scholar to identify SNPs in putative genes. A combination of keywords such as- “aspirin”, “NSAID”, “pharmacogenetics”, “polymorphisms”, “SNPs”, “gene variant”, “colorectal adenoma” and “colorectal cancer”, was used to search for relevant literature published between 1990 and 2015 in the search engines. On PubMed, this resulted in 103 abstracts. All studies presenting original data on SNP and colorectal adenoma or carcinoma risk association, or interaction between SNP and aspirin (or NSAID) use in relation to colorectal adenoma or carcinoma were retrieved and reviewed. Types of articles reviewed included original article, editorial letter, conference abstract, observational study, meta-analysis, review, systemic review and randomized controlled trial. The UK-CCSG study collaborators crosschecked candidate SNP selection from the literature. From the published studies reviewed until May 2015, a total of 42 SNPs from 15 candidate genes were selected for analysis .ALOX15, IL16, MDR1, MGST1, NFkB, UGT1A6, PTGS1 and PTGS2) [CES2 and PAFAH1B2 were selected as these genes have been shown to metabolize aspirin in intestine and blood, respectively, but had not been tested for interaction with aspirin use in relation to CRC risk previously [CYP2C9) [IkBkB, CCAT2, 20p12 locus, NCF4 and ODC1) [We selected genes and SNPs as follows: (i) from previous studies of colorectal adenoma or cancer where statistical evidence–after correction for multiple testing–of an interaction with NSAID use (including aspirin) had been observed (d PTGS2) –23; (ii)eviously , 25; (ii[CYP2C9)  and; (ivnd ODC1) , 16, 27.No SNPs were genotyped for the NIH-CCFR dataset as these samples had been the subject of genome-wide genotyping  using IlP<0.001 after multiple tests correction). Overall, 15 of the 42 SNPs were removed from analysis as the observed MAF was <4%. For the remaining SNPs, all were in HWE and the MAF between the two datasets was similar  of <4%; or there was evidence of being out of Hardy Weinberg equilibrium (HWE) in controls ( similar .2) in controls in both datasets  and 95% confidence interval (CI) for associations between known epidemiological risk factors, “aspirin only” use, and SNP genotype with CRC risk were assessed using logistic regression models adjusted for age, sex and study site separately for each study. Each SNP was tested for association with colorectal cancer risk using logistic regression and assuming an additive model of inheritance. The interaction of SNP genotype with “aspirin only” use in relation to CRC risk was investigated. Participants who took aspirin only were included in the analysis, whereas, users of both aspirin and other NSAIDs  or other NSAIDs only  were excluded from the analysis. Regular aspirin use was defined as daily aspirin intake for 3 months or longer in the UK-CCSG dataset whereas it was defined as twice a week aspirin intake for more than a month in the NIH-CCFR dataset . GxE intP-values were two-sided and the significance threshold was set at ≤0.001 to allow for multiple testing . All analyses were conducted in Stata V12 . Checklist outlining information about the justification for the study and the methodology employed is provided in Meta-analysis of the association and interaction tests for the SNPs and “aspirin only” use with CRC risk in both datasets was carried out. Estimates of log odds ratios and standard error from both datasets, which were adjusted for age, sex and study site, were used to calculate combined risk estimates using random-effects models where the log OR were weighted by the method of DerSimonian and Laird . To testOverall, the UK-CCSG dataset consisted of 1910 cases and 1275 controls and the NIH-CCFR dataset consisted of 1415 cases and 986 controls. Characteristics of the study population are presented in P = 0.001, I2 = 0), rs2302615  and rs11694911  at CCAT2 and the ODC1 gene locus, respectively  and ODC1 rs2302615 variant allele    Tables. UGT1A6 gene showed evidence of an interaction between aspirin-only use and CRC risk , whereby aspirin users with wild-type genotype were associated with 23% lower CRC risk  compared to negligible risk reduction in variant allele carriers   with rs2070959 (R2 = 0.90), showed similar evidence   Tables. evidence  Figs. NoPinteraction = 0.006): aspirin users with the wild-type genotype showed a 26% lower risk of colon cancer than non-users with wild-type genotype , whereas, no difference in risk reduction was observed in variant allele carriers regardless of the aspirin use status (Pinteraction = 0.008). No evidence of interaction was observed between either SNP, aspirin use and rectal cancer risk  whereas the NIH-CCFR cohort is enhanced for early onset (47.7% of cases below the age of 50 years). Since aspirin is prescribed prophylactically for cardiovascular diseases after the age of 50, it is likely that the participants under the age of 50 may not be exposed to aspirin. Hence, we performed sensitivity analysis by excluding all participants under the age of 50 and carried out GxE interaction test between only aspirin use, SNP variant allele and CRC risk. Compared to the analysis carried out in all participants, we observed similar direction and magnitude of interaction between SNPs in the CCAT2 gene locus with reduced CRC risk, which was noteworthy after multiple test correction and congruent with the literature [CYP2C9 and colon cancer risk was complemented by the assessment of the difference in CYP2C9 protein expression between colon and rectum in the Human Protein Atlas. Medium protein expression is observed in the colon but not rectal tissue [In the current meta-analysis, we observed association between the SNP at terature , 32. Potl tissue , suggestl tissue . As CYP2l tissue . Hence, ODC1’s rs2302615 A allele. Whilst the association was not significant after multiple test correction, it was complemented by the study carried out by Zell et al. which observed decreased probability of survival in patients diagnosed with rectal cancer and carrying rs2032615 A allele  but not in colon cancer patients  [ODC1 is carried out by E-box transcription factors, chiefly- transcriptional activator c-MYC and transcriptional repression MAD1. Tissue specific RNA-seq data in the Human Protein Atlas show 7.5 and 5.8 transcripts per million of MAD1 in colon and rectum tissues, suggesting slightly higher expression of MAD1 in the colon compared to rectum [ODC1 promoter containing A allele in two colon cancer cell lines- HT29 and HCT116 [Interestingly, we observed a potential site-specific association for decrease in colon cancer risk in the presence of 85–3.63) . Ornithio rectum . Coupledd HCT116 , it is lUGT1A6 gene gave suggestive evidence of decreased risk of CRC or colon cancer but not in variant allele carriers. The UDP glucuronosyltransferase 1A6 (UGT1A6) protein is involved in metabolism of salicylic acid through glucuronidation [UGT1A6 SNPs, aspirin use and colorectal adenoma risk has been observed whereby, carriers of variant allele were associated with lower risk of colorectal adenoma in aspirin users compared to non-users [UGT1A6 SNP variant allele, any NSAID (including aspirin) use and CRC risk in the current study revealed no difference in the direction of association compared to the interaction involving aspirin only use (data not presented). Whilst testing for the source of the contradiction between these observations was beyond the remit of the current study, it is plausible to attribute the differences to the distinction between the key genetic and epigenetic molecular factors between adenoma and carcinoma that may make carcinoma cells more sensitive to aspirin intervention [UGT1A6. As the wild-type allele has lower metabolic activity compared to the variant allele, participants with wild-type genotype are likely to retain active metabolites of aspirin longer than the mutant allele carriers; hence, deriving greater chemopreventive benefit. Moreover, this is in keeping with observations in chemoprevention trials CAPP1 and CAPP2 where the chemopreventive effect of aspirin was restricted to colorectal cancer incidence with apparently limited impact on colorectal adenoma development [UGT1A6 SNPs with colon cancer has been observed previously, albeit without statistical significance due to small study size consisting of only 422 cases and 481 population controls [Upon assessing GxE interaction, aspirin-only users with wild-type genotype of the SNPs in nidation . A studynidation , 38, henon-users , 39. Thirvention . One of rvention , 42. Oneelopment , 44. Addcontrols .ODC1 gene has previously been shown to be associated with increased risk of colorectal adenoma [MGST1 locus, we found decreased CRC risk among aspirin users; this finding is consistent in direction but not as strong as the findings from the GWAS study of Nan et al. 2015 where the study involved over 8000 cases and controls. However, for individuals carrying the variant allele, that study showed significantly increased risk of CRC among NSAID users, a result inconsistent with our observations. The current study didn’t observe evidence of interaction between variant alleles of SNPs in previously reported genes- ALOX15, IL16, MDR1, NFkB, PTGS1 and PTGS2- with aspirin only use and CRC or colon cancer risk. This could in part be attributed to the differences in the choice of drug definition for carrying out the interaction test. Previous studies carried out GxE interaction test with any NSAID use (including aspirin) and CRC risk. The broad range of NSAIDs used to carry out the test includes drugs with different pharmacokinetic and pharmacodynamics mechanisms thus precluding the assessment of the true source of interaction.Interestingly, SNP rs11694911, which is located downstream to the ad hoc power calculation was performed in Quanto V1.2.4 (http://biostats.usc.edu/Quanto.html) using the outcomes observed in the current study to assess the number of cases and controls needed to carry out an interaction test between UGT1A6 SNP, aspirin only use and colorectal cancer risk with two-sided P value of 0.002 (corrected for multiple tests) and 80% power. Assuming an unmatched case-control (1:1 ratio) study design, population CRC risk (Po) of 6%, SNP minor allele frequency of 31% with additive mode of inheritance, genetic effect (RG) of 5%, environmental effect  of 26% and population prevalence of aspirin users (PE) of 20%: we would require 7057 cases and controls to observe an GxE interaction effect (RGE) of 20%. This highlights that the current study with a combined total of 3325 cases and 2262 controls had limited statistical power for GxE interaction. Whilst we did not adjust analysis for multiple test correction, hence increasing the probability of false positive results, we did observe results for some of the SNPs that are congruent with previously published studies where correction for multiple testing was applied. Even though these results are consistent with ours, all positive results should be regarded as hypothesis-generating and should be investigated in other independent datasets.An Additionally, we could not investigate whether there was a dose-response relationship among aspirin users because dose information was not recorded in the NIH-CCFR dataset. Other NSAIDs were also used by participants in both studies but numbers precluded detailed examination of each. Compared with previous studies where analysis for interaction was carried out in all users of NSAIDs–and thus limiting identification of the true source of interaction as each NSAID affect different biological pathways–our study was restricted to interaction in aspirin-only users.CYP2C9, ODC1 and UGT1A6, may be involved in the modification of CRC or colon cancer risk independently or in conjunction with aspirin use.In summary, our results suggest variant alleles of the genes involved in aspirin pathways, S1 File(XLSX)Click here for additional data file.S2 File(DOCX)Click here for additional data file.S3 File(DOCX)Click here for additional data file.S1 Table+ Smoking includes cigarettes, cigar and pipes. ^Alcohol includes beer, cider, wine, sherry, other fortified wine, sake, champagne and spirits.(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+ Observed MAF in controls of the two datasets were compared using Fisher’s exact test. A rs16973225 and rs5277 were only genotyped in cases in the NIH-CCFR dataset.* Observed minor allele frequency (MAF) in controls was compared to the MAF reported for Phase I GBR and Phase I CEU population from 1000 Genomes database in UK-CCSG and NIH-CCFR datasets respectively using Fisher’s exact test. (DOCX)Click here for additional data file.S5 Tablea, Total number of subjects calculated from only CCO, UQM and FHCRC study sites.P-value calculated using Fisher’s exact test for categorical variables and Mann-Whitney U test for continuous variables.b, LTRI = Lunenfeld-Tanenbaum Research Institute, Ontario; USC = University of Southern California; UoM = University of Melbourne; MC = Mayo Clinic; FHCRC = Fred Hutchinson Cancer Research Centre; UHI = University of Hawaii Cancer Center.(DOCX)Click here for additional data file.S6 TableP-value for association adjusted for age, sex and study site.*CI, Confidence Intervaln, Number of subjects.(DOCX)Click here for additional data file.S7 TableP-value for association adjusted for age, sex and study site.*P-value for Cochran’s Q-test for heterogeneity.+CI, Confidence Interval.(DOCX)Click here for additional data file.S8 TableP-value for association adjusted for age, sex and study site.+CI, Confidence Intervaln, Number of subjects.(DOCX)Click here for additional data file.S9 TableP-value for association adjusted for age, sex and study site.*P-value for Cochran’s Q-test for heterogeneity.+CI, Confidence Interval.(DOCX)Click here for additional data file.S10 TableP-value for association adjusted for age, sex and study site.*P-value for Cochran’s Q-test for heterogeneity.+CI, Confidence Interval.(DOCX)Click here for additional data file.S11 TableP-value for association adjusted for age, sex and study site.+P-value for interaction between SNP variant allele, aspirin use and colorectal cancer risk calculated using Likelihood ratio test. P-value is adjusted for age, sex and study site.*OR, Odds RatioCI, Confidence Intervaln, Number of subjects.(DOCX)Click here for additional data file.S12 TableP-value for association adjusted for age, sex and study site.+P-value for interaction between SNP variant allele, aspirin use and colorectal cancer risk calculated using Likelihood ratio test. P-value is adjusted for age, sex and study site.*P-value for Cochran’s Q-test for heterogeneity~OR, Odds RatioCI, Confidence Interval.(DOCX)Click here for additional data file.S13 TableP-value for association adjusted for age, sex and study site.+P-value for interaction between SNP variant allele, aspirin use and colorectal cancer risk calculated using Likelihood ratio test. P-value is adjusted for age, sex and study site.*P-value for Cochran’s Q-test for heterogeneity~OR, Odds RatioCI, Confidence Interval.(DOCX)Click here for additional data file.S1 Fig(DOCX)Click here for additional data file.S2 Fig(DOCX)Click here for additional data file.


Some observational studies have examined the association between dietary whole grain intake and all-cause mortality, but the results were inconclusive. We therefore conducted a meta-analysis to summarize the evidence from cohort studies regarding the association between whole grain intake and all-cause mortality. Pertinent studies were identified by searching PubMed, Embase and Web of Knowledge, up to February 28, 2016. Study-specific estimates were combined using random-effects models. Eleven prospective cohort studies involving 101,282 deaths and 843,749 participants were included in this meta-analysis. The pooled relative risk of all-cause mortality for the highest category of whole grain intake versus lowest category was 0.82 . There was a 7% reduction in risk associated with each 1 serving/day increase in whole grain intake . No publication bias was found. This analysis indicates that higher intake of whole grain is associated with a reduced risk of all-cause mortality. The findings support current recommendations for increasing whole grain consumption to promote health and overall longevity. The number of deaths was 52.8 million globally in 2010 and the deaths from non-communicable diseases rose from just under 8 million in 1990 to 34.5 million in 2010, accounting for two of every three deaths [Grains, also called cereals, are the seeds of grasses cultivated for food. Whole grains are the entire seed of a plant and this seed  is made up of three key edible parts- bran, germ, and endosperm . The braAccumulating epidemiological evidence indicates that high intake of whole grains might decrease the risks of obesity/abdominal fatness , type 2 n = 1) [n = 2) [n = 1) [A flow chart for the search process is presented in Figure n = 1) ; 2) newe [n = 2) , 36; and [n = 1) . One stu [n = 1)  that was [n = 1) –32.n = 7) [n = 4) [Eleven prospective cohort studies from ten publications –32 publin = 7) , 30, 32  [n = 4) , 28, 31. [n = 4)  to 367,4I2 = 76.6%, Pheterogeneity < 0.001) , Egger's test (P = 0.575).Ten studies from nine publications , 28–32 w) Figure , Begg's I2 = 92.4%, Pheterogeneity < 0.001) , Egger's test (P = 0.895).Ten studies from nine publications –32 with ) Figure , Begg's I2 = 0%, Pheterogeneity = 0.558) compared with the pooled relative risk for the association of dietary whole grain intake with all-cause mortality among population in America. For studies in which with adjustment for BMI, physical activity level, alcohol consumption, diabetes, blood pressure, or serum cholesterol level, the pooled relative risk was higher than that in studies without adjustment. However, meta-regression analysis showed that no variables may account for the heterogeneity across studies on the association between whole grain intake and total mortality.In subgroup analyses of the association between whole grain intake and all-cause mortality, all estimates, despite disparate magnitudes, showed inverse associations, and all were statistically significant except in male population Table . MoreoveIn sensitivity analyses, we sequentially excluded one study at a time and reanalyzed the remaining data. The pooled relative risks for sensitivity analyses were still statistically significant and similar to the overall estimate. The pooled relative risks for all-cause mortality ranged from 0.81  when the Health Professionals Follow-up Study by Wu et al.  was exclTo our knowledge, the present study is the first meta-analysis quantitatively assessing the association between dietary whole grain intake and all-cause mortality. In this meta-analysis, increased whole grain intake was significantly associated with a reduced risk of total death. The risk of all-cause mortality was decreased by 18% for individuals in the highest category intake of whole grain, compared with it in the lowest. The results of the dose-response meta-analysis suggested that each additional 1 serving whole grain intake daily may lower the risk of all-cause mortality by 7%.The protective effect of whole grain consumption on risk of all-cause mortality is biologically plausible. Whole grains are rich source of fiber, vitamins, minerals, phenolic compounds, phytoestrogens, and other phytochemicals, which may potentially explain whole grains' favorable effects , 37–42. In subgroup analyses, among male population, no significant inverse association was found. This might be due to that the number of cohorts in the group was only two, so the limited sample sizes had inadequate test power to examine the association. Most studies included in the present meta-analysis adjusted for some of the major potential confounders or risk factors, such as BMI, physical activity level, alcohol consumption. When we stratified studies by adjustment for specific confounders, the magnitudes of the association in the subgroups with adjustment were considered to be more reliable because these factors may be confounders in the analysis due to their associations with both the risk of all-cause mortality and whole grain consumption. When we restricted the analysis to studies that were adjusted for diabetes, blood pressure, or serum cholesterol level, the pooled relative risk was more close to 1 than the overall estimate. Dietary whole grain intake could influence the risk of all-cause mortality via several different mechanisms. Controlling for any of the intermediate variables, such as diabetes, blood pressure, or serum cholesterol level, in the hypothesized casual pathway between dietary whole grain and all-cause mortality might lead to overadjustment and thus bias the result towards null . TherefoAlthough most studies indicated inverse association between whole grain consumption and all-cause mortality, significant heterogeneity was observed in this meta-analysis, which could limit the interpretability of the pooled estimate. There were several potential explanations for the observed between-study heterogeneity. First, the way in which these articles compared categories was different. Several studies calculated the estimation for the highest quintile vs. the lowest , 30, 32,The present meta-analysis had some strengths. To the best of our knowledge, this was the first meta-analysis to explore the relationship between dietary whole grain intake and all-cause mortality. A highlight of this present meta-analysis was the prospective design of the included cohort studies, which should have greatly reduced the potential selection and recall bias. The large study population  enabled us to provide sufficient statistical power to quantitatively assess the relationship between whole grain intake and all-cause mortality. Most of the studies included in the meta-analysis adjusted for large number of major confounders. Because of this, the pooled estimate may be less susceptible to confounding bias.However, some limitations of the meta-analysis should be considered. First, it was difficult to consistently define and precisely evaluate the intake of whole grains or whole grain products in epidemiological studies and some degree of measurement error was inevitable, and reporting whole grain intakes as the actual amount of whole grain intake per dry weight had been pointed out . Most stIn conclusion, the present meta-analysis indicates that an increased whole grain intake is associated with a reduced risk of all-cause mortality. These findings add to and extend the evidence that an increased dietary whole grain intake may exert healthy effects and decrease the risk of all-cause mortality and support current recommendations to increase whole grain consumption to promote health and overall longevity.We followed standard criteria for reporting meta-analyses of observational studies . We perfPublished studies were included if they 1) reported all-cause/total mortality as the outcome of interest; 2) were conducted in a general population; 3) used prospective cohort design; 4) presented information on whole grain intake as the exposure of interest; and 5) provided estimates of relative risk (RR) or hazard ratio (HR) with confidence intervals (CIs) or standard errors or the data necessary to calculate these estimates.The studies were excluded if they 1) were non-English language; 2) were not original articles, such as reviews, letters, comments, etc.; and 3) had repetitive data on the same population. We used the most recent report or report with the largest number of cases if there were duplicates.Data abstracted from each study were as follows: first author's name, year of publication, country in which the study was conducted, duration of follow-up, gender of the study population, age group of the subjects, sample size of the cohort, number of deaths, assessment tool used to measure dietary whole grains, type of exposure, categories of dietary whole grains and relative risks and 95% confidence intervals for all-cause mortality associated with those categories, and covariates included for adjustment in multivariable models as well.We calculated the pooled relative risk of all-cause mortality and its 95% confidence interval for the highest category of dietary whole grain intake versus the lowest using DerSimonian and Laird random-effects model, which incorporate both within- and between-study variations .For the dose-response meta-analysis, numbers of deaths and persons/person-years for at least 3 whole grain intake categories and the mean or median values of the categories were needed. If the mean or median values were not reported in the studies, the estimated midpoints of the categories were used for substitution. When the highest categories were open-ended, we assumed that the open-ended categories were of the same amplitude as the adjacent categories .We used To assess the source of variations among studies, we carried out subgroup analyses stratified by gender, study location, follow-up period, and cohort size. We also conducted analyses stratified by whether studies adjusted for potential important confounders, including total energy intake, BMI, smoking status, alcohol consumption, physical activity level, or adjusted for potential intermediate variables, including diabetes, blood pressure, and serum cholesterol level. For each stratification variable, heterogeneity between subgroups was evaluated by conducting meta-regression analyses.Q and I2 statistics, and results were defined as heterogeneous for a P value < 0.10 or I2 > 50% [We conducted sensitivity analyses to assess the influence of individual studies by excluding one study at a time and created a sensitivity plot. Publication bias was assessed using Begg's test  and EggeI2 > 50% .P value < 0.05 was considered statistically significant unless otherwise specified.All statistical analyses were conducted using Stata/SE software, version 12.0 . Two-sided


CETP C-629A polymorphism with CHD risk and lipid changes. Four public databases were searched, and data from 17 qualified articles were extracted in duplicate and analyzed by STATA software. Overall association of C-629A with CHD risk was nonsignificant in 5441 patients and 7967 controls. Subgroup analyses by ethnicity revealed significance only in Caucasians, with the odds of CHD being 1.18, 1.43 and 1.41 under allelic, genotypic and dominant models, respectively (P < 0.001). Similarly, the -629C allele increased the corresponding risk of myocardial infarction by 1.23-, 1.28- and 1.29-fold (P < 0.02). The association of C-629A with CHD was significantly strengthened in prospective and large studies. Moreover, carriers of the -629C allele had significant higher levels of circulating CETP , but lower levels of high-density lipoprotein cholesterol (HDL-C)  relative to the -629AA homozygotes. The probability of publication bias was low. Our meta-analytic findings collectively demonstrate that the -629C allele was significantly associated with an increased risk of CHD in Caucasians, and this association may be mediated by its phenotypic regulation on circulating CETP and HDL-C.Lipid metabolism plays an essential role in the pathogenesis of atherosclerosis, a major cause for coronary heart disease (CHD). Cholesteryl ester transfer protein (CETP) is an important glycoprotein involved in lipid metabolism by transferring cholesteryl esters to apolipoprotein B-containing lipoproteins in exchange for triglycerides. The objective of this meta-analysis was to evaluate the association of  A clear understanding of how CETP genetic loci regulate lipid metabolism associated with CHD is therefore a challengeable task. To take a step further, we in this study meta-analytically evaluated the association of a promoter functional polymorphism, C-629A (rs1800775) in CETP with CHD risk and lipid changes. This polymorphism was reported to be a Sp1/Sp3 transcription factor binding site that can regulate the transcriptional activity of human CETP promoter [It is widely accepted that lipid metabolism plays an essential role in the pathogenesis of atherosclerosis, a major cause for coronary heart disease (CHD) , 2. The  studies . However rs180077 in CETP CETP C-629A polymorphism and CHD risk [CETP C-629A polymorphism and circulating lipid changes [Figure CHD risk –18. Ther changes , 19–24.Table P = 0.053) and gender composition was comparable (P = 0.111). Mean levels of BMI (P = 0.005), smoking status (P < 0.001), hypertension (P = 0.001) and diabetes (P = 0.182) were significantly higher in CHD patients than in controls. By contrast, controls had significant higher levels of circulating HDL-C (P = 0.001) and Apo-AI (P = 0.026) than patients.CHD patients were slightly older than controls , 15%  and 14%  increased risk under allelic , homozygous genotypic (-629CC genotype versus -629AA genotype) and dominant (-629CC genotype plus -629AC genotype versus -629AA genotype) models, respectively. These associations were obsessed by moderate heterogeneity, with the corresponding I2 statistic of being 69.1%, 50.2% and 60.4%. There was a low probability of publication bias except for dominant model (Egger's test: P = 0.055)  with moderate heterogeneity (I2 = 53.0%).Table P < 0.001 for all), without observable heterogeneity . In contrast, the effect estimates were in an opposite direction, albeit nonsignificant in Middle Easterns across three genetic models.Stratifying study groups by ethnicity identified significance only in Caucasians, with the odds of CHD being 1.18, 1.43 and 1.41 respectively under allelic, homozygous genotypic and dominant models (P = 0.015), homozygous genotypic (P = 0.015) and dominant (P = 0.001) models, with borderline heterogeneity. By source of controls, the effect estimates were roughly comparable between hospital- and population-based studies, with significant heterogeneity.In subgroup analyses by CHD subtypes, the -629C allele was observed to significantly increase risk of myocardial infarction by 1.23-, 1.28- and 1.29-fold respectively under allelic (P < 0.05) without evident heterogeneity, while this risk was reduced towards the unity in retrospective studies.By study design, the -629C allele seemed to confer a 21% to 28% increased risk for CHD in prospective studies across three genetic models (P = 0.004), homozygous genotypic  and dominant  models with borderline heterogeneity, while an opposite yet nonsignificant association was identified in the small studies (< 500 subjects).When the analysis was restricted to the large study (≥ 500 subjects), pooled risk estimates were statistically significant under allelic  or -629C allele  relative to the -629AA homozygotes, with evident heterogeneity. By contrast, there was a reduced yet nonsignificant trend in circulating triglycerides for -629CC genotype or -629C allele carriers, and the probabilities of heterogeneity and publication bias were low.Table P = 0.003) or -629C allele  when compared with the -629AA homozygotes, with moderate heterogeneity and a low probability of publication bias. The -629CC genotype or -629C allele was associated with higher circulating LDL-C than the -629AA genotype, while no significance was reached. Similarly, the -629CC genotype or -629C allele was associated with lower Apo-AI but higher Apo-B than the -629AA genotype, with no observable heterogeneity.Circulating HDL-C was significantly reduced in carriers of the -629CC genotype .To further seek possible causes of clinical heterogeneity, meta-regression analyses that modeled age, male gender, BMI, smoking, dyslipidemia, hypertension, diabetes, circulating triglycerides, total cholesterol, HDL-C, LDL-C, CETP, Apo-AI and Apo-B if available under study were conducted, and none of these factors contributed significantly to the association of Cumulative analyses by ascending publication years indicated no substantive change in the direction of effect estimates with the addition of subsequent studies . In addiCETP C-629A polymorphism with the risk of CHD and lipid changes by summarizing data from 17 articles. The key finding of this study was that the -629C allele was significantly associated with an increased risk of CHD in Caucasians, and this association may be mediated by its phenotypic regulation on circulating CETP and HDL-C. The importance of the current study lies in deepening our understanding of the functional aspects of CETP genetic variation involved in the pathogenesis of CHD.The objective of this meta-analysis was to evaluate the association of CETP gene TaqIB (rs708272) polymorphism as an instrument has demonstrated that circulating CETP may play a causal role in the pathophysiology of CHD [CETP genetic loci associated with CHD risk and responsible for the changes of biologically relevant lipids. The conduct of this meta-analysis therefore represents a supplement to medical research and deepens our understanding of the genetics of CHD.Lately, a large comprehensive meta-analysis choosing y of CHD , althougy of CHD , such asCETP C-629A mutation can alter susceptibility to CHD in Caucasian populations, at least in part, through its phenotypic regulation on circulating CETP and HDL-C. Several cautionary notes regarding the interpretation and extrapolation of this finding should be sounded. First, genetic heterogeneity across ethnicities is a common phenomenon gripping a majority of association studies. It is of interest to found that CETP -629C allele was significantly associated with an increased risk of CHD only in Caucasians, while this association was reversed to be protective in Asian and Middle Eastern populations. As a matter of fact, linkage disequilibrium patterns are generally believed to be diverse across races or ethnicities. For example, the linkage of a genetic variant with another functional variant was usually strong in one ethnic group but weak or nonexistent in another [CETP genetic alterations with increased large cholesterol-enriched HDL particles [CETP C-629A polymorphism with CHD risk had a biological basis, as this polymorphism also accounted for the changes of circulating CETP concentrations. In addition, this association was strengthened after restricting analysis to the prospective and large studies, which further verifies the robustness of our meta-analytic findings.As indicated in this meta-analysis,  another . Second,articles . Moreovearticles –32. In aarticles  and Niacarticles  that raiCETP was meta-analyzed, which is clearly not sufficient to support the contributory role of CETP in the pathogenesis of CHD and lipid regulation, as other polymorphisms in or flanking CETP might synergize or antagonize the impact of C-629A. Fourth, the association of CETP C-629A polymorphism with CHD risk and circulating lipid changes is not based on the same dataset due to limited number of qualified studies. Fifth, as with most meta-analyses, publication bias might be possible because only published articles were retrieved and the ‘grey’ literature (articles in languages other than English) was not reviewed. In view of these limitations, the jury must refrain from jumping at a conclusion until further verification of our findings in large, long-term, well-designed prospective studies.A number of possible limitations should be recognized for this meta-analysis. First, only summary data from published papers were abstracted, and it could yield further insights if individual participant data were analyzed. Second, we were unable to glean various potential confounders  from all eligible studies, and only five studies provided complete confounding data. We adopted meta-regression analyses in an attempt to account for this limitation, while no significance was identified. Third, only one promoter polymorphism, C-629A in CETP C-629A polymorphism underlying CHD, future studies on the relationship between CETP genetic defects and CHD susceptibility need to focus on gene-to-environment interactions, especially on the impact of the -629C allele on circulating lipid changes.Taken together, our meta-analytic findings demonstrate that the -629C allele was significantly associated with an increased risk of CHD in Caucasians, and this association may be mediated by its phenotypic regulation on circulating CETP and HDL-C. Although further investigations are required to elucidate the molecular mechanisms of This meta-analysis of observational studies was conducted according to the PRISMA  statement . All obsPubMed/Medline, Embase, Scopus and Google Scholar electronic resources were searched on July 21, 2016 to seek articles of potential relevance, by using subject headings  and (‘polymorphism’ or ‘variant’ or ‘variation’ or ‘mutation’ or ‘SNP’). The bibliographies of retrieved articles were also reviewed for articles that might be missed.The eligibility of each article was justified by two investigators (Shouwei Lin and Rong Lin) by reading the title and abstract, and if necessary the full text. Meanwhile, the period and location, if available, for study subjects collected were recorded to judge whether there were multiple publications from the same study. If so, the publication with a larger sample size was retained.CETP C-629A polymorphism with CHD risk or circulating lipid changes was evaluated; (3) the absolute counts of C-629A genotypes between CHD patients and controls or the circulating lipid concentrations across C-629A genotypes were provided. In addition, articles were not taken into account if they were conference abstracts/posters, case reports, editorials and narrative/systematic reviews.To be more specific, three inclusion criteria were proposed: (1) only English-language publications were considered; (2) the association of CETP C-629A polymorphism between CHD patients and controls or mean (standard deviation) concentrations of circulating CETP, triglycerides, HDL-C, LDL-C, Apo-AI and Apo-B across C-629A genotypes, as well as age, male gender, body mass index (BMI), smoking, dyslipidemia, hypertension, diabetes, circulating triglycerides, total cholesterol, HDL-C, LDL-C, CETP, Apo-AI and Apo-B, if available, between CHD patients and controls. For the sake of consistency, circulating triglycerides, total cholesterol, HDL-C, LDL-C, Apo-AI and Apo-B were expressed in mg/dL and CETP in μg/mL.The same two investigators (Shouwei Lin and Rong Lin) independently extracted data from each qualified article according to a jointly formulated protocol, including first author's surname, publication year, ethnicity, diagnostic criteria of CHD including coronary stenosis or myocardial infarction, study design (retrospective or prospective design), source of controls , matched situation, sample size, absolute genotype counts of I2) statistic (range: 0% to 100%). Statistical heterogeneity was reported to be significant if the I2 statistic is over 50%, which is a generally accepted cutoff value [Unadjusted odds ratio (OR) and weighted mean difference (WMD), along with 95% confidence interval (95% CI) were calculated by using a random-effects model with the DerSimonian & Laird method to pool individual effect-size estimates under all circumstances. The magnitude of statistical heterogeneity across studies was represented by the inconsistency index .The assessment of publication bias was made by the Begg's funnel plots and Egger's asymmetry tests. The Egger's test can inspect funnel plot asymmetry by determining whether the intercept deviates significantly from zero when regressing the standardized effect estimates against their precision . P < 0.1


Few studies have reported the relationship between retinal artery occlusion (RAO) and plasma homocysteine (Hcy) levels. Our goal was to evaluate the association between the plasma Hcy level and the risk of RAO disease. Several databases were searched for all published studies that involved Hcy and RAO. Six studies evaluated hyperhomocysteinemia (hHcy) in retinal artery occlusion patients and controls; the incidence of hHcy in patients with RAO was higher than the control and the pooled odds ratio (OR) was 6.64 : 3.42, 12.89). Subgroup analyses showed that the ORs were 4.77  in Western countries, 22.19  in Asian countries, 9.70  in the age matched group, 11.41  in the sex matched group, 9.70  in the healthy control group, and 6.82  in the sample size >30. The mean plasma Hcy level from 5 case-control studies was higher than controls, and the weighted mean difference (WMD) was 6.54 . Retinal artery occlusion is associated with elevated plasma Hcy levels. Our study results suggest that hHcy is probably an independent risk factor for RAO. RAO is the blocking of trunks or branches of the retinal circulation, which leads to the ischemic infarction of the affected retinal tissue7. Central retinal artery occlusion (CRAO) always produces severe visual loss9, and branch of the retinal artery or cilio-retinal artery occlusion provokes a remarkable visual impairment and damage to the visual field10. Isolated visual symptoms mean that patients may first be treated by ophthalmologists, rather than directly visiting an emergency department. When patients present with RAO, ophthalmologists should ensure that they receive appropriate acute and secondary preventive treatment11 because it is an important warning of cardiovascular and cerebrovascular events15. This means that patients with incidents of RAO are at increased risk of cardiovascular disease and ischaemic stroke18. Although there are many factors, such as embolism, haemorrhage, and intraluminal thrombosis that have been suggested to play a major role in the primary mechanisms of coronary and cerebral vascular disease and RAO in common, they do not fully explain the pathogenesis of RAO. Over the last few years, Hcy has aroused a considerable amount of interest for physicians. Elevated plasma Hcy levels appeared to be clearly associated with an increased risk of cardiovascular and cerebrovascular events23, which suggests that Hcy plays a role in the pathogenesis of vascular disease, similarly in retinal artery occlusion. Recently, the association between Hcy and the risk of RAO has been reported in some studies32, but the conclusion is contradictory and inconclusive, partially due to the relatively small sample size of individual studies, sampling effects and lack of statistical power. To overcome such limitations, we conducted a meta-analysis of the published case–control studies that directly evaluated and compared Hcy and RAO.Retinal artery occlusion (RAO) is a cause of ocular morbidity that mainly affects patients older than 60 yearsWe initially searched 88 unique papers: PubMed (n = 39), Embase (n = 49) and Cochrane Library (n = 0), 2 additional publications were produced through a review of article citations. Sixty-seven studies were eliminated as follows: duplicats n = 19), case reports (n = 22), letters (n = 6), reviews (n = 5), irrelevant (n = 9), and conference papers (n = 6). The full text of 23 potentially included papers were reviewed; we excluded three publications that did not have controls, two publications that were ocular fundus disease controls, one paper that did not have clear controls, seven papers did not have complete data, and two papers that had duplicated data. Finally, eight case-control studies were involved in our meta-analysis. The flow diagram showing the study selection is presented in Fig. 9, case r34. The studies were published between 2000 and 2015. These studies were performed in the United States, Europe, and Asia. The controls were healthy and hospital based individuals without RAO were included. The sample size of the studies ranged from 8 to 105, and the total number of cases was 307 in the hHcy group and 196 in the concentration of plasma Hcy group, and the numbers of controls were 542 and 350, respectively. The OR varied from 2.53 to 85.5, and the WMD ranged from 1.9 μM to 11.99 μM. All NOS scores were greater than 5, showing that the methodological quality was good.The main information of the eight case control studies is shown in Table 34 shows the relationship of hHcy and the risk of RAO using the random effects model of combining OR. Six studies, including 307 RAO patients and 542 controls, have assessed the proportion of hHcy34. The rate of hHcy was significantly greater than the controls . Substantial heterogeneity was observed .Figure 32 shows the results of the random-effects models of combining WMD for the average plasma Hcy level of the RAO patients and controls in 5 studies32, and a total of 196 cases and 350 controls were included in the meta analysis. The results showed that the plasma Hcy level is higher than the controls . It indicates that the increased plasma Hcy might be an independent risk factor for RAO. There was evidence of heterogeneity for this outcome .Figure P = 0.05, I2 = 73%), while little heterogeneity was found in the European and American region . The pooled ORs were 4.77  in the European and American Region and 22.19  in the Asian Region respectively , while insignificant in the age-matching subgroup . According to the RAO inclusion number, it was divided into two subgroups. Two trails26 that enrolled patients less than 30 were excluded, but this did not change the overall risk estimate (OR: 6.82 [95% CI: 4.19 to 11.10]); no evidence of heterogeneity was observed among the remaining studies , as shown in Fig. 34. Heterogeneity was still present through subgroup analysis; the overall combined OR was not substantially altered. Furthermore, the exclusion of a single study and its effect on the overall OR estimate was evaluated. The combined OR ranged from 5.45  to 8.15 , and each study did not significantly affect the summary estimate, which suggested the high stability of the result. The heterogeneity was reduced when study24 or26 was eliminated. The above analyses consistently illustrate that there was a significantly positive association between hHcy and RAO.There was significant heterogeneity among the six studies concerning the relation between hHcy and RAO; thus, we conducted subgroup analysis and sensitivity analysis. The region, sex, age, the selection of controls, and the number of included RAO patients may influence the summary combination, so we performed subgroup analysis for these factors. According to region, the studies were divided into two subgroups, and significant heterogeneity was found in the Asian region  while excluding three studies29 that included RAO patients less than 30. We also performed sensitivity analysis by omitting one study each time, and the combined WMD ranged from 5.28 μM  to 7.84 μM . This yielded a similar result with similar heterogeneity.Substantial heterogeneity was observed among the 5 studies, which showed the association between plasma Hcy and RAO. Thus, a sensitivity analysis was performed to explore the potential source. It exhibited lower results with lower heterogeneity  was calculated. The Nfs values were 24.44 in the hHcy group and 136.10 in the plasma Hcy group. The Nfs was bigger than the number of observed studies included in the meta-analyses, which implied a non-significant publication bias.Visual inspection of the Begg funnel plot did not identify substantial asymmetry Fig. . The fai35. The risk factors of thrombophilia include hyperlipidemia, arteriosclerosis, homocysteine and so on37. However, the association between homocysteine and the risk of RAO has not been fully understood, and consistent conclusions have not been provided. Some articles have suggested that higher plasma Hcy levels were associated with the risk of RAO26, while some reports indicated that there is no significant difference among them27. Therefore, we performed a meta-analysis to provide a quantitative assessment of the relationship. Our meta-analysis of eight case-control studies demonstrates that hHcy or higher plasma Hcy concentrations will increase the risk of RAO. Moreover, in an attempt to produce more stable, robust and credible results, we performed a subgroup analysis and a sensitivity analysis, and the results did not materially change the pooled results. Therefore, our results displayed that the elevated plasma level of Hcy or hHcy is an independent risk factor for RAO. The estimation of plasma Hcy levels should be considered in patients with RAO. The identification of elevated plasma Hcy levels may also reduce the patient’s risk of developing a cardiovascular event because retinal emboli are an important predictor of stroke and other vascular diseases38.Thrombophilia is a complex hypercoagulable state that increases the risk of thrombosis. It plays a significant pathoetiologic role in the development of RAO, which is associated with stroke, coronary artery disease, atrial fibrillation, and carotid stenosis39, Hcy significantly increased inducible nitric oxide synthase and endothelial nitric oxide synthase and decreased glutathione levels indicating oxidative-nitrosative stress. On the other hand, the increased expression of matrix metalloproteinase (MMP) and the decreased expression of tissue inhibitor of metalloproteinase (TIMP) and tight junction proteins (zonula occluden-1) indicate vascular remodelling. Hcy promots the expression of profibrogenic cytokines and influences the MMP/TIMP balance40. The MMP is known to regulate the activities of collagenase and gelatinase41. MMP degrades collagen, allows for smooth muscle cell migration within a vessel, and induces vascular remodeling42.The underlying mechanisms may explain why Hcy can increase the risk of RAO. On the one hand, Hcy-induced endothelial cell dysfunction, endothelial cell apoptosis and proliferation disorders were exacerbated2 > 50%). Several factors might explain the heterogeneity, such as the characteristics of populations, the source of populations, and the adjustment of confounding factors. Our analysis suggests that two studies26 with small sample sizes probably contribute to the heterogeneity and the heterogeneity will dispear (I2 = 0) after delete these two articles .It was not surprising that there was substantial heterogeneity among the publications in this meta-analysis  group guidelinesWe searched the following databases: PubMed, EMBASE through the OVID platform, and CoChrance Library. No languages or publication year were restricted; the final search was carried out in April, 2016. Literature retrieval was conducted to identify all relevant trials that compared the association between Hcy and RAO. The following terms were used as key words or Mesh words: “homocysteine” OR “hyperhomocysteinemia” and “retinal artery occlusion” OR “retinal artery thrombosis”. In addition, we searched manually on the reference lists of retrieved articles and recent reviews for additional relevant articles to identify possible relevant articles.Studies that were considered eligible in this meta-analysis met the following criteria: (1) The study design was a case-control study. (2) The interesting exposure was plasma Hcy level or hHcy, and the interesting outcome was RAO. (3) Plasma Hcy concentrations or ORs and their 95% CIs or data were used to calculate them. (4) If multiple publications for the same study were available, we used the most recent study. If necessary, we used data from the early publications. (5) The control was healthy or hospital-based patients with other eye disease. (6) The control had no ocular fundus disease.(1) Editorials, letters to the editor, case reports, previous meta-analyses, review articles, and meeting abstracts. (2) The control group was not clearly defined. (3) Incomplete raw data.46. The NOS uses a “star” rating system to judge methodological quality. The scores range from 0 to 9, with scores 0–4 indicating low quality and scores 5–9 indicating high quality.Two authors (Xuetao Huang and Yezhen Yang) independently conducted the study selection and data extraction; any disagreements were resolved by discussion, and the authors reached a consensus on all details. The following information was extracted from each article: first authors, year of publication, country, age and sex, the number of cases and controls, the Hcy levels in the cases and control subjects, the number of hHcy in the cases and controls, the OR of the association between hHcy and RAO. We assessed the quality of each study using the Newcastle-Ottawa Scale (NOS)Because the characteristics of populations, the diagnosis of RAO, the source of controls and adjusting confound factors varied among all studies, we further conducted a subgroup analysis and sensitivity analysis to explore the possible source of heterogeneity. We also assessed the stability of the results by removing a single study each time to recalculate the pooled data.We used Review Manager 5.2.6 to calculate the pooled data. The odds ratios (ORs) were used as a common measure to evaluate the association between hHcy and the risk of RAO. The weighted mean differences (WMDs) were used to compare the plasma Hcy concentrations between the case and control subjects. The pooled data were calculated by using a random-effects model to achieve a more conservative assessment.2 statistics. For the Q statistic, P < 0.1 was considered statistically significant heterogeneity among the combined studies. I2 values of 50% or more indicate a significant heterogeneity.Statistical heterogeneity was estimated using Cochran’s Q test and the Ifs to assess the presence of publication bias, if the Nfs is bigger, the results are more stable, and the results are more credible. If the value of Nfs for one comparison turned out to be smaller than the number of included studies, it showed a significant publication bias. The Nfs is calculated by the formula Nfs = [ΣZ(pi)]2/1.6452 − n, where n equals the number of observed trials, and p = 0.05.Publication bias was qualitatively assessed by the visual inspection of funnel plot asymmetry of the logarithmic OR against their standard errors and the MD against their standard errors. We also used N


The association between abdominal obesity (as measured by waist circumference (WC) and waist-to-hip ratio (WHR)) and colorectal cancer (CRC) has not been fully quantified, and the magnitude of CRC risk associated with abdominal obesity is still unclear. A meta-analysis of prospective studies was performed to elucidate the CRC risk associated with abdominal obesity. Pubmed and Embase were searched for studies assessing the association between abdominal obesity and CRC risk. Relative risks (RRs) with 95% confidence intervals (95% CIs) were pooled using random-effects model of meta-analysis. Nineteen prospective cohort studies from eighteen publications were included in this meta-analysis. A total of 12,837 CRC cases were identified among 1,343,560 participants. Greater WC and WHR were significantly associated with increased risk of total colorectal cancer , colon cancer , and rectal cancer . Subgroup analyses further identified the robustness of the association above. No obvious risk of publication bias was observed. In summary, abdominal obesity may play an important role in the development of CRC. Colorectal cancer (CRC) is a major public health concern, as it is one of the leading causes of cancer deaths in the Western world . AlthougThe prevalence of overweight and obesity is increasing dramatically in most parts of the world, and can lead to obesity-related cancers, including postmenopausal breast cancer, colorectal, endometrial, esophagus, kidney, lung, pancreatic, thyroid, and gallbladder cancers –14. HoweThe risk of CRC in obese individuals, especially those with higher abdominal obesity has not been fully quantified, and it is also unclear whether abdominal obesity is an independent risk factor of CRC. Therefore, a comprehensive systematic review and meta-analysis of prospective studies was performed to estimate the risk of CRC associated with abdominal obesity.The present study was planned, conducted, and reported in adherence to the “Meta-Analysis of Observational Studies in Epidemiology (MOOSE)” guidelines . We systStudies were included if they met all of the following criteria: (a) prospective cohort studies; (b) the exposure of interest was abdominal obesity (measured using WC and/or WHR); (c) the outcome of interest was colorectal cancer; (d) risk estimates of colorectal cancer associated with abdominal obesity were available, such as relative risks (RRs) or hazard ratios (HRs) with 95% confidence intervals (CIs); and (e) the risk estimates were adjusted for other confounding factors. Studies were excluded if they focused on colorectal mortality or recurrence.Using a standardized extraction form, the following data were extracted from each study: the first author’s last name, publication year, country, study period, age range, sex, number of cases, number of participants, data collection, measures of abdominal adiposity, cancer sites, most fully adjusted risk estimates with their corresponding 95% CIs for each category of abdominal adiposity measures, and adjustment for potential confounding factors. Risk estimates reported by gender were extracted separately. Newcastle–Ottawa Scale (NOS) was used to assess the quality of included studies . AccordiQ and I2 statistics. For the Q statistic, P<0.1 was considered statistically significant; and for the I2 statistic, I2 more than 40% indicated substantial heterogeneity across included studies, and metaregression was further utilized to find the source of heterogeneity. Both Begg’s and Egger’s tests were performed to investigate potential publication bias [P-value of less than 0.05 was considered to be statistically significant.HR was directly considered as RR across the present study. The pooled RRs with 95% CIs were calculated using a random-effects model with the method of DerSimonian and Laird . Becauseion bias ,23. To eA total of 714 articles were initially identified from Pubmed and Embase databases. After removing the duplicated articles, 382 studies were included for further assessment. Of these articles, 355 studies were excluded after reading the titles and the abstracts. After full-text review of the remaining 27 articles, 18 studies ,18,24–38The general characteristics of the included studies are presented in I2 = 40.1%, P=0.015), while low heterogeneity was observed for colon cancer . On the other hand, no evidence of heterogeneity for rectal cancer . Subgroup analyses further identified the robustness of the association between abdominal obesity and CRC risk , as well as for colon cancer , while no heterogeneity was observed for rectal cancer . Subgroup analyses further identified the robustness of the association between abdominal obesity and CRC risk  To capture all relevant information, studies were included after a comprehensive, systematic search of the literature by a multidisciplinary team including specialists in gastrointestinal endoscopy, gastroenterology, and clinical epidemiology and using a broad search strategy. (b) Most of the included studies adjusted for nearly all the important covariates including age, education, family history of colorectal cancer, physical activity, alcohol intake, smoking etc. (c) The present meta-analysis is based on prospective studies, so we have effectively avoided recall and selection bias.in vivo and in vitro experiment as a pleiotropic hormone being mitogenic, anti-apoptotic, pro-angiogenic, and proinflammatory in various cellular systems [Obesity is considered one of the important risk factors for many types of cancer, especially for CRC. However, the mechanisms that might underlie the association between excess weight and CRC remain unclear. Several mechanisms have been proposed to explain how general and central obesity enhances colorectal neoplasm risk. Previous studies have demonstrated that the fat itself can also influence CRC risk . Adipocy systems . The rel systems . In addi systems ,46. IGF  systems , which m systems . The inv systems .Available epidemiologic evidence suggests that abdominal obesity (as reflected by high WC and WHR) may be more predictive of colon cancer risk than overall obesity (high BMI) –53. ThisHowever, several limitations in this meta-analysis should be considered. First, most of the included studies did not provide the risk estimates controlling for weight change during follow-up, and they could not exclude the impact of weight change during follow-up on the association between abdominal obesity and CRC. Second, although individual studies have considered a wide range of potential confounders in their analyses, we cannot fully exclude unknown or residual confounding factors which may have influence on our findings. Third, although our analysis indicate that both higher WC and WHR increase the risk of colorectal cancer, colon cancer, and rectal cancer, few studies have conducted further adjustments between abdominal obesity measurement and BMI to try to clarify their respective roles. Finally, as with any meta-analysis, publication bias is a matter of concern, because small studies with null results tend not to be published. Although there was no evidence of publication bias, we cannot exclude such bias because of low statistical power due to limited number of studies.In summary, findings from this meta-analysis of prospective studies provide evidence that abdominal obesity may play an important role in the development of colorectal cancer. This positive association also exists in both men and women, different geographic region, and different anatomical site. Further large prospective studies are necessary to evaluate whether the association between central obesity and CRC is biased by BMI.