Increasing evidence connects non-alcoholic fatty liver disease (NAFLD) to cardiovascular disease (CVD). The aim of this study is to assess whether and to what extent the excess risk of CVD is conferred by NAFLD in a meta-analysis. We systematically searched PubMed, EmBase, Web of Science, and Cochrane Library for reports published between 1965 and July 3, 2015. Studies that reported data on association between NAFLD and adverse cardiovascular events or mortality were included. Thirty-four studies  were included. NAFLD was not associated with overall mortality  and CVD mortality . However, NAFLD was associated with an increased risk of prevalent  and incident  CVD. For some specific CVDs, NAFLD was associated with an increased risk of prevalent  and incident  coronary artery disease (CAD), prevalent  and incident  hypertension, and prevalent  atherosclerosis. In conclusion, the presence of NAFLD is associated with an increased risk of major adverse cardiovascular events, although it is not related to mortality from all causes or CVD. Non-alcoholic fatty liver disease (NAFLD) is the leading cause of chronic liver disease worldwide, with a prevalence as high as 30% in the general population256Although CVD mortality rates have decreased by as much as 50% in several developed countries since the 1970s, it still remains the most frequent cause of death worldwideOur initial search yielded 1641 records, of which 1560 remained after removal of duplicates . Thirty-91011121314151617181920212223242526272830313233343536373839404191011121314151617192022232425262932333637384041182127283031343591011121314151617182123242526282930333436373940192022273132I2 statistic for heterogeneity between studies was 65.4%, with P value for the χ2 test 0.008, suggesting substantial between-study heterogeneity. In the subgroup analyses that had more than one comparison, individuals with NAFLD had significantly higher risk of overall mortality than those without NAFLD in studies using hospital-based study design and with relatively low study quality  . The I2  quality . Sensiti quality .I 2 = 64.9%, P = 0.002). In the subgroup analyses, the pooled HRs did not differ significantly by most of the study-level factors except for studies with hospital-based design (P = 0.032), study design (P = 0.007), follow-up duration (P = 0.079), BMI (P = 0.001), and presence of diabetes (P = 0.054) were significant at the level of 0.1, and these five covariates were entered into the multivariate meta-regression analysis. After including these five covariates in the model, the τ2 changed from 0.0539 to 0.01191, which means that 77.90% of heterogeneity between the studies can be explained by these covariates. Sensitivity analysis indicated that the non-significant difference in CVD mortality was not materially changed in the leave-one-out analyses by omitting one study in turn, with pooled HRs range from 1.02 (95% CI: 0.86–1.21) to 1.16 (95% CI: 0.87–1.55)  . Potentid design . Univari87–1.55) .I 2 = 79.8%, P < 0.001). Subgroup analyses indicated that the significantly higher risk was not seen participants with mean age over 50 years, in studies including exclusively Asian participants, with relatively low study quality, and not adjusting for age or BMI/obesity or smoking  . There w smoking . Sensiti smoking .I 2 = 55.1%, P = 0.038). The pooled HRs did not differ significantly in the subgroup analyses that had more than one comparison  . Potentimparison . The leamparison .I 2 = 80.2%, P < 0.001). In the subgroup analyses, the higher risk was not significant in studies using population-based design and including exclusively diabetic participants (P = 0.035), male percent (P = 0.061), study design (P = 0.092), ethnicity (P = 0.078), and study quality (P = 0.013) were significant at the level of 0.1, and these five covariates were entered into the multivariate meta-regression analysis. After including these five covariates in the model, the τ2 changed from 0.06891 to 0.01922, which means that 72.10% of heterogeneity between the studies can be explained by these covariates. Sensitivity analysis indicated that the significantly higher risk was not materially changed in the leave-one-out analyses by omitting one study in turn, with pooled ORs range from 1.49 (95% CI: 1.28–1.75) to 2.02 (95% CI: 1.52–2.69)  . There wicipants . In the 52–2.69) .Only one cohort study was available for assessment of association between NAFLD and incident CAD. The HR reported by the study was 2.31 (95% CI: 1.46–3.65) .I 2 = 0.0%, P = 0.525). The magnitude and direction of the associations were unaltered across studies in the subgroup analyses  . There wanalyses . The leaanalyses .I 2 = 55.9%, P = 0.059). In the subgroup analyses, the higher risk was not significant in studies not adjusting for age or BMI/obesity or smoking  . Potenti smoking . Sensiti smoking .I 2 = 34.0%, P = 0.218). In the subgroup analyses, the higher risk was not significant in studies including exclusively non-Asian participants and not adjusting for age or BMI/obesity or smoking  . There w smoking . Sensiti smoking . = 1.37, 95% CI: 0.86–2.19, I 2 = 86.4%, P < 0.001) and CVD mortality  but significantly increased the incident CVD risk   . Subgrou< 0.001) . Sensiti< 0.001) .P = 0.020) indicated there was potential publication bias when assessing NAFLD and CVD incidence in cohort studies  (P = 0.013) and Begg-Mazumdar test (P = 0.012) indicated there was potential publication bias when assessing NAFLD and CAD prevalence in cross-sectional studies   . Both Eg studies . After u04–1.77) .The main results of our meta-analysis are the following: (1) NAFLD was not associated with overall mortality and CVD mortality; (2) NAFLD was associated with an increased prevalence and/or incidence of other adverse cardiovascular events, including CVD, CAD, hypertension, and atherosclerosis; (3) NASH was not associated with overall mortality and CVD mortality but was associated with an increased incidence of CVD. These results are important given the high prevalence of NAFLD in the general population and the concerns raised by the adverse metabolic profile associated with this disease and NASH.Based on the current evidence from the literature, the association between NAFLD and adverse cardiovascular events is mixed. Several researchers reported that NAFLD is associated with CVD in diabetic patients2433428In our study, we found no association between NAFLD or NASH and deaths from all causes or CVD. This result was similar with a cohort study from Sweden. This study followed up a cohort of 144 patients with NAFLD for over 13.7 years, and found patients with NAFLD had similar survival to the general Swedish population (matched for age and gender). However, their study found the risk of death was increased in patients with NASH46In addition, our study indicated that NAFLD was an independent risk factor in determining cardiovascular events  by pooling the multiple-adjusted data together, and NASH was an independent risk factor in determining CVD. Hamaguchi and colleagues53et al.45Our study showed that NASH, a more severe stage of NAFLD, was associated with higher risks in overall mortality, CVD mortality, and incident CVD with larger HRs, compared with NAFLD, although the risks were not significant in overall mortality and CVD mortality. A previous study showed that NASH is associated with a more severe inflammatory and insulin-resistant state that promotes atherosclerosiset al.Recently, Targher This meta-analysis provides the most definitive and convincing evidence so far of NAFLD-related risk of cardiovascular events. The findings were robust and applicable across a broad range of populations. However, the study has several limitations. First, the most pooled analyses revealed heterogeneity among studies. Although subgroup analysis and meta-regression analysis gave some clues to explain the heterogeneity, these are unlikely to have fully accounted for heterogeneity. Therefore, the results of the meta-analysis must be interpreted with caution. Because of potential additional heterogeneity in the populations, designs, and analyses of the various studies, we assumed that the true effect being estimated would vary between studies, in addition to the usual sampling variation in the estimates. To account for the heterogeneity, we used the random-effect model to combine the results of the primary studies. The random-effect approach provides some allowance for heterogeneity in studies beyond sampling error. This does not necessarily rule out the effect of heterogeneity between the studies, but one can expect a very limited influence because of it. Second, the meta-analysis is based on observational studies, which leaves the possibility that residual confounding factors, including measurement errors, affect the relation between NAFLD and adverse cardiovascular events. Third, data were only available for analysis for risks of overall mortality, CVD mortality, prevalent/incident CVD, prevalent/incident CAD, prevalent/incident hypertension, and prevalent atherosclerosis, and other adverse cardiovascular events are not analyzed because studies on other adverse cardiovascular events are quite few for solid meta-analysis. More studies are needed to confirm the relationship between NAFLD and risk of other specific cardiovascular diseases. Fourth, there were limited studies in several subgroup analyses, which may lead to low statistical power in these analyses. Another important limitation of these data, which warrants further investigation by future studies, was that a standard definition of diabetes was not used across studies and ascertainment of diabetes probably varied between studies. Additionally, although we did subgroup analysis by the presence of diabetes, data on exclusively diabetic participants and non-diabetic participants were quite few. Further studies are needed to examine the potential contribution that the presence of diabetes might have on the excess risk of adverse cardiovascular events.In conclusion, our analysis shows that the presence of NAFLD is not associated with overall mortality and CVD mortality but is associated with an increased risk of other adverse cardiovascular events, including CVD, CAD, hypertension, and atherosclerosis. Future studies should evaluate strategies and interventions to prevent cardiovascular disease progression in individuals with NAFLD.We followed the PRISMA guidelinesWe included studies that met the following criteria: participants aged 18 years or older; cross-sectional design, prospective design, or retrospective design; the association between NAFLD and adverse cardiovascular events or mortality was assessed; and reported data on odds ratios (ORs) or hazard ratios (HRs) with confidence intervals (CIs) or sufficient information to calculate these, for the association between NAFLD and adverse cardiovascular events or mortality. Studies were excluded if they did not provide information to calculate the point estimate, did not make comparison between NAFLD and adverse cardiovascular events or mortality, or were review studies. Articles that clearly did not meet inclusion criteria were rejected on initial review. If uncertainty existed, the full text of the article was reviewed. Two reviewers (FW and JH) independently assessed all potentially relevant studies for inclusion. Disagreements were resolved by consensus.Study characteristics were extracted independently by two researchers (PM and YH). We extracted risk estimates (95% CI) for different genders and disease severities separately when possible. If a study reported more than one measure of adverse cardiovascular events, each adverse cardiovascular event was extracted separately. The most adjusted estimate was included when a study reported more than one risk estimate. The quality of each study was assessed by two researchers (FW and PM), using the Newcastle-Ottawa Scale (NOS) recommended by Wells and colleaguesvs non-NAFLD patients reported in each study, using non-NAFLD patients as the reference group. The impact of NAFLD histological subtypes (NASH) on adverse cardiovascular events was also examined.Associations with continuous outcome variables were expressed as weighted mean differences (WMDs) with 95% CIs. For evaluation of the relative risk of adverse cardiovascular events, the effect size was estimated as ORs or HRs with 95% CIs, according to NAFLD patients I-squared (I 2) statistic and Q-statistic were used to explore the heterogeneity among studies. Large I 2 (>50%) or P < 0.1 for Q-statistic suggests substantial heterogeneity among studies. We separately analyzed cross-sectional and longitudinal studies. Moreover, the results of studies defining NAFLD by ultrasound, liver biopsy, CT images, or liver enzyme elevation are presented separately. We did several subgroup analyses: study design , mean age of the participants (≥50 years or <50 years), ethnicity (non-Asian or Asian), presence of diabetes (diabetic participants or non-diabetic participants or combined), study quality (high vs relatively low), and adjustment of some major risk factors (yes vs no). When eight or more comparisons were available, the effect of several variables including age, publication year, male percent, study design, follow-up duration (for cohort studies), body mass index (BMI), ethnicity, smoke, presence of diabetes, and study quality was assessed by meta-regression analysis to further evaluate the source and strength of heterogeneity. We also performed sensitivity analyses by removing each individual study from the meta-analysisWe used the random-effect model in this meta-analysis to take into account heterogeneity among studies, because the study design and measuring time were different across studiesHow to cite this article: Wu, S. et al. Association of non-alcoholic fatty liver disease with major adverse cardiovascular events: A systematic review and meta-analysis. Sci. Rep.6, 33386; doi: 10.1038/srep33386 (2016).


Polymorphisms in NKX2-3 gene have been inconsistently associated with Crohn's disease (CD) and ulcerative colitis (UC). To generate large-scale evidence on whether NKX2-3 polymorphisms are associated with CD or UC susceptibility we have conducted a meta-analysis of 17 studies involving 17329 patients and 18029 controls. A significantly increased CD or UC risk was observed in persons carrying a G allele at rs10883365 polymorphism (A/G) compared with those with a A allele. . In the subgroup analysis, a significantly increased CD risk was found in both Europeans and Asians. For rs11190140 polymorphism (C/T) and CD risk, the risk estimate for the allele contrast was OR = 1.201 (1.136–1.269). This meta-analysis provided a robust result that persons with a G or T allele may have a moderately increased risk of CD, and suggested that rs10883365 polymorphism was also a candidate gene polymorphism for UC susceptibility. Inflammatory bowel diseases (IBDs) are chronic inflammatory disorders characterized by chronic relapsing inflammation of the gastrointestinal tract that affect 0.1% of Western populations, comprising two major forms, Crohn's disease (CD) and ulcerative colitis (UC)34678910NKX2-3, located on 10q24, is a member of a family of genes that encodes transcription factors containing homeodomains and, therefore, is implicated in basic developmental functions. During development, NKX2-3 is expressed in midgut and hindgut mesoderm and spleen, as well as in pharyngeal endoderm121517181520Thus, a quantitative synthesis may help to provide clearer evidence on the association of such genetic polymorphisms with IBD. In the present study, we conducted a meta-analysis of all eligible studies to quantitatively assess the associations between three common polymorphisms (rs10883365 and rs11190140) in the NKX2-3 gene and IBD susceptibility.The combined search yielded 75 references, of which 31 were duplicate studies, 9 were reviews, 4 were about cell studies, 8 were only with abstracts, 7 reported other mutations, 1 reported other disease. Finally, a total of 15 articles were finally included. Among them, one publication1115161718192021222324252627The summary of meta-analysis for the NKX2-3 polymorphisms with CD is shown in A total of 5484 patients and 4863 controls were investigated for rs11190140 variant, a significant association was found in all genetic models. see  Similar Seven studies with 4996 UC patients and 5479 controls for rs10883365 polymorphism were investigated. Meta-analysis findings of associations between rs10883365 in NKX2-3 gene and the risk of UC were shown in Sensitivity analysis showed no single study qualitatively changed the pooled ORs. see  MoreoverFunnel plots and Egger's test were performed to assess publication bias. No publication bias was detected for rs10883365 polymorphism. . Similarly, no publication bias was detected for T vs C contrast of rs11190140 polymorphism in CD. see  As shownPresently the mechanisms of the etiology and progression of IBD are far from clear. Several genes have been identified to be associated with IBD risk, including NOD2, NKX2-3 and IL-23. Recently, genome-wide association studies (GWAS) have identified SNPs implicating hundreds of replicated loci for common traits and becomes a powerful tool to detect the susceptibility genes in the IBD diseases16293132A meta-analysis can combine results from individual studies to overcome the limitation of small sample sizes and inadequate statistical power, and produce a single estimate of the major effect. Recently, accumulated meta-analysis has been performed to investigate the association of genetic variants with susceptibility to CD or UC. Polymorphisms in several genes, including ATG16L1 T300ANKX2-3, located on chromosome 10q24, is predominantly expressed in mesoderm of midgut and hindgut during embryonic mouse development452 > 0.8). The results indicated that rs10883365 and rs11190140 are in perfect LD (r2 = 1.0).To the best of our knowledge, the present study involving 37039 subjects represents the first comprehensive meta-analysis investigating the association between NKX2-3 polymorphisms (rs10883365 and rs11190140) and IBD susceptibility. For the analysis of rs10883365 polymorphism, a significantly increased CD risk was observed in all genetic models. In the ethnicity-stratified analyses, significant association was found both in Asian and European populations. Similar results were found between the rs11190140 variant and risk of CD where a significant association was found European population (no Asian population reported). No between-study heterogeneity was observed in most genetic models. Sensitivity analysis indicated that when excluding studies departed from HWE, the pooled OR still did not change, demonstrating the results of this meta-analysis were stable. Since these two SNPs are close to each other, we use 1000 Genomes Pilot sequence data to identify whether these SNPs are in linkage disequilibrium (LD) . In addition, the reference lists of reviews and retrieved articles were checked by hand-search for additional potential studies. A study reported results from more than one population was considered as separate studies.Studies were considered eligible if they had to meet the following criteria: (1) association between NKX2-3 polymorphisms and risk of IBD (CD or UC), (2) case-control or cohort studies. Studies were excluded for the following reasons: (1) articles only with an abstract and review articles (2) no control population, (3) studies considered overlapped with other studies.Two authors extracted the following data independently from each of the eligible articles: first author, publication year, ethnicity of the participants involved (categorized as Europeans or Asians), number of cases and controls, information  of cases and controls, and number of genotypes or allele frequency in cases and controls. Study authors were contacted for detailed data when there was insufficient information to determine the relationship between genetic polymorphism and IBD risk. Disagreements were resolved by discussion between the two authors.Pooled odds ratios (ORs) and 95% confidence intervals (CIs) used to assess the strength of the association between IBD risk (CD or UC) and NKX2-3 polymorphisms (rs10883365 or rs11190140). The significance of the pooled OR was determined by the Z-test; a P-value of <0.05 was considered significant. The Hardy-Weinberg equilibrium (HWE) in the control group was assessed, and a P < 0.05 was considered as significant disequilibrium. For rs10883365 polymorphism, the pooled ORs were estimated for G versus A, GG versus AA, GA versus AA, dominant model (GG + GA versus AA), and recessive model (GG versus GA + AA). Because of only three studies1949Between-study heterogeneity was evaluated by using the Chi-square based Q test and I2 test52Sensitivity analysis was carried out to evaluate the stability of the results after sequential removal of each study or by excluding those studies deviated from HWE. In addition, cumulative meta-analyses were carried out for each polymorphism through assortment of studies with publication time. Graphical evaluation of funnel plots and Egger's linear regression test were performed to assess publication biassupplementary figures


MIF gene has been associated with numerous diseases, such as arthritis and cancer. However, most of the reports concerning the association of MIF with these and other pathologies are inconsistent and remain quite controversial. Therefore, we performed a meta-analysis from 96 case-control studies on -173 G/C MIF SNP and stratified the data according to the subjects geographic localization or the disease pathophysiology, in order to determine a more meaningful significance to this SNP. The polymorphism was strongly associated with an increased risk in autoimmune-inflammatory, infectious and age-related diseases on the dominant  and the recessive models . Also, significant association was found in the geographic localization setting for Asia, Europe and Latin America subdivisions in the dominant  and overdominant models . Afterwards, we implemented a network meta-analysis to compare the association of the polymorphism for two different subdivisions. We found a stronger association for autoimmune than for age-related or autoimmune-inflammatory diseases, and stronger association for infectious than for autoimmune-inflammatory diseases. We report for the first time a meta-analysis of rs755622 polymorphism with a variety of stratified diseases and populations. The study reveals a strong association of the polymorphism with autoimmune and infectious diseases. These results may help direct future research on MIF-173 G/C in diseases in which the relation is clearer and thus assist the search for more plausible applications.Human macrophage migration inhibitory factor (MIF) is a cytokine that plays a role in several metabolic and inflammatory processes. Single nucleotide polymorphism (SNP) -173 G/C (rs755622) on  Human macrophage migration inhibitory factor (MIF) is a cytokine involved in several metabolic and inflammatory processes, that have been widely studied in recent years. MIF is a versatile protein that acts as a potent upstream regulator of the immune system. This 12.5-kDa protein also plays an important role as a cytokine, chemokine, and even as an enzyme. As a cytokine, MIF has proinflammatory activities through the induction of the expression of other inflammatory cytokines and countering the effects of glucocorticoids  in the MIF promoter sequence, which is the most widely studied variant in this gene. This polymorphism has been associated with the pathophysiology of arthritis, infections and inflammatory-related diseases   Studies reporting the frequency of polymorphism rs755622, and (2) studies presenting both patients and control group data.The exclusion criteria were: (1) Studies with genotype frequency data not included or incomplete, (2) reports that had self-contradictory data, and (3) studies with control data not in Hardy-Weinberg equilibrium.Only articles meeting the inclusion criteria were included. Different studies from the same authors were compared to discard data duplications, and when a duplicate was observed or highly suspected, only the data from the earlier report was considered. Partially duplicated studies where not excluded from the meta-analysis when the duplication was restricted to the controls, as the comparison made in the meta-analysis is patients vs. controls and not controls vs. controls. However, these studies were not considered for the geographical distribution studies.The data involved patients and healthy individual control groups, and the data from each single study were considered to be a cohort, a group of individuals sharing a defining characteristic. When two or more subtypes of a disease were reported, the data were merged into a single cohort. On the other hand, when two or more unrelated diseases were reported in the same study or the data were clearly divided between two or more countries, data were managed as two different cohorts.Cohorts were grouped in subdivisions according to four different settings, all subdivisions include at least five different cohorts Table . The geop2, 2p(1-p), and (1-p)2, respectively. The Chi-square test was estimated considering one degree of freedom and a P < 0.5.Deviation from HWE for each study was evaluated to improve the design and quality of the analysis. The proportions of genotypes CC, CG and GG were considered to be et al  statistics test. The pooled effect size (OR and SMD) was assessed based in the random-effect model if heterogeneity was considered statistically significant . If not, then the fixed-effect model was used. To evaluate the specific effects of ethnicity and study quality, we also performed a subgroup analysis of different specific effects.We calculated the heterogeneity using χhttp://www.metafor-project.org/doku.php/metafor. Holm-Bonferroni correction was performed following the standard procedure  was performed according to Bucher et al. , in whicA total of 96 publications were considered for this analysis, comprising 100 different cohorts for the rs755622 SNP. The general information and data mined from each study are provided in Table The hierarchical clustering represented in heat-maps Figure  indicateP < 0.0001, P-value Het: < 0.0001), recessive , and homozygous  models.The meta-analysis results are listed in Tables P-value: 0.0437, P-value Het: < 0.0001) and homozygous models  and the latter in the dominant model alone . IBD subdivision was not significantly associated to the polymorphism in any subdivision.Other diseases, such as infectious and autoimmune-inflammatory diseases, were significantly associated with the polymorphism in all of the genetic models, except for the allelic model. In the disease setting, only arthritis and cancer presented a significant association, the former in the dominant , ,  and overdominant models , , .In the geographic localization setting, Africa did not present a significant association in any subdivision, while Asia, Europe and Latin America presented a significant association in the dominant (Asia OR: 0.76 CI:0.69–0.84, Forest plots for representative subdivisions and genetic models are presented in Figure Representative Funnel plots are presented in Figure The Funnel and forest plots corresponding to the entire meta-analysis in the five different genetic models are presented in Figure A NMA was performed using a simple method proposed by Bucher et al.  that invWe presented the frequencies of all of the studies included in the meta-analysis for both controls and patients, divided according to geographical setting Figure , Table 5MIF-173 G/C polymorphism with a variety of diseases were included. Considering their significant diversity, these studies were categorized according to the determined common characteristics, such as the geographic localization or pathophysiology of the disease, instead of the classical comparison of overall results. Clustering analysis for the geographic setting showed a tendency to cluster studies of African, European, Latin American, and North American subdivisions. The Asian and Middle East subdivisions showed no cluster tendency, probably due to the admixture of the studies, nevertheless, we decided to maintain the complete setting in further analyses, in order to compare among different regions.Genetic association studies are important for epidemiological analyses, as these studies can identify candidate genome regions that contribute or associate to specific diseases Lee, . In the After the analysis of the different genetic models used in the meta-analysis, the allelic model showed less significant associations with the subdivisions, reflecting not only a lack of difference between controls and patients at allelic level but also a higher heterogeneity at the genotype level. We included this model in the analysis to determine if a strong influence of a single allele on disease was present, however, this was not the case. The localization of the polymorphism on the promoter sequence indicates that its biological significance is related to differences in the transcript expression levels (Matia-García et al., MIF polymorphisms on IBDs has previously been studied in other meta-analyses, including four published studies focused on Asian and European cohorts, and the results showed a clear association between the MIF -173 GC SNP and IBD (Hao et al., MIF -173 GC SNP is associated with an increased cancer risk, obtaining heterogeneous results. For instance, one study in Asian population reported a significant relation between this polymorphism and the risk of developing gastrointestinal cancer or hematological malignancy (Tong et al., Arthritis and cancer, two conditions previously associated with MIF and the -173 GC SNP, presented significant associations in at least one model, while IBD did not demonstrate any association. In this context, the influence of The relation of the polymorphism with arthritis has also been reported in meta-analyses, and a significant relation was observed, but the included studies were almost exclusively limited to European populations (Lee et al., In the present study, we observed an association between GC and CC genotypes and increased disease risk in the Asian, European and Latin American subdivisions, according to the results in the dominant and overdominant models, which were also supported by the results of the European subdivision in the recessive model and Asian subdivision in the homozygous model. These results indicate the importance of geographic localization and ethnic origin in disease even in the same country. For example, in the United States, the highest incidence and mortality of colorectal cancer is observed in non-Hispanic black people, while the lowest incidence is observed in the Asian-American/Pacific islander group (Siegel et al., We also compared the genotypic and allelic frequencies obtained in each geographic localization subdivision in the present meta-analysis with previously reported results. The allele frequencies of the African, Latin American, Asian and European subdivisions were consistent with those reported in the 1,000 genome project (Consortium, MIF polymorphism with these diseases (Harper et al., Another setting used to categorize the data was based in disease pathophysiology, that includes a subdivision composed of conditions with degenerative components or associated to age-related diseases. Considering the preponderant role of MIF in the immune response and in aging, we also included subdivisions composed of diseases characterized by a strong and pathologic inflammatory response, a pathogenic autoimmune process, or both. The results of the present meta-analysis showed a clear relation of the polymorphism with disease in this subdivision in the dominant, recessive and homozygous models, correlated with several studies exploring the role of MIF in homeostasis and inflammation, and since the pathogenesis of most age-related diseases led to a special type of chronic inflammation, the so called “inflammaging,” is therefore not surprising that we observed an association of the MIF microsatellite polymorphism rs5844572, suggest the existence of a selective pressures acting on the MIF locus. In the case of rs5844572, that is also localized in MIF promoter sequence, the pressure of continuous exposition to infectious diseases would explain the high prevalence of low-expression MIF alleles in Africa. This proposition may not be applicable to the -173 GC polymorphism, considering that the frequency of MIF low-expression allele G is the same in patients and in controls, even though both are lower than the reported by the 1,000 genome project (Table Infectious subdivision presented an association with the polymorphism in all genetic models but the allelic. The association was unexpected, considering that MIF has an effect in the expression of TLR4, the signal transducing molecule of the LPS receptor (Roger et al., Autoimmune-inflammatory and inflammatory subdivisions also presented significant associations indicating an increased risk for the GC and CC genotypes according to multiple models. Similarly, the autoimmune subdivision presented an increased risk for CC genotype carriers according to the recessive and homozygous models.MIF -173 GC SNP on the disease risk for different subdivisions. In this case, we compared the association of the polymorphism with diseases in each subdivision and model of the pathophysiology setting, and the results indicate whether one subdivision has a greater effect on disease risk. To our knowledge, this study is the first to perform a NMA in a genetic association model. Our results indicate that in the recessive model, the autoimmune subdivision has a stronger association with the polymorphism than the age-related and autoimmune-inflammatory subdivisions. On the other hand, results indicate a stronger association of the infectious subdivision than the autoimmune-inflammatory subdivision. Thus, the polymorphism and its effect in the expression of the protein, have a significant role in the genesis or development of the diseases in the autoimmune and infectious subdivisions. Also, it indicates that even when age-related and autoimmune-inflammatory diseases may be associated to the polymorphism of MIF, this association is weaker, and less likely to be of biological interest.All of the subdivisions in the pathophysiology setting presented an increased risk for C allele carrier genotypes. Bucher indirect comparison is an NMA statistical method used to determine which of two different treatments is more efficient, comparing the magnitude of the treatment effect. Thus, we performed a simple NMA (Bucher et al., After the Holm-Bonferroni correction was applied some subdivisions ceased to be significantly associated with disease. Arthritis subdivision lost significance in all models, while autoimmune-inflammatory subdivision was still significantly related to disease only in the overdominant model, and the infectious subdivision maintained it in the dominant and allelic models. It should be noted that subdivisions of the geographical localization setting majorly maintained its significant association with disease in different models, indicating a strong correlation with disease. These results should be considered carefully, as the Bonferroni test is highly conservative and stringent. Even though arthritis subdivision presented no significant association according to this correction, the relation between the diseases included in the subdivision remains true, as all of them are rheumatoid disorders, and may not be treated as different traits. Also, the test has limitations regarding the geographical localization setting, considering that different countries does not necessarily have different populations, and one country may have many.P-values is conservative and stringent, however, a stringent validation was needed, considering that the study proposes the inclusion of different traits in the same analysis. Nevertheless, the present study a meta-analysis including case-control studies of the relation of a wide variety of diseases with a polymorphism. Further experiments are required to confirm the relations found and detail the subyacent mechanisms, and this may help direct future research on MIF-173 G/C in diseases in which the relation is clearer and thus assist the search for more plausible applications.In summary, our results indicated that the GC and CC genotypes are associated with arthritis, cancer, and importantly, all of the subdivisions proposed in the pathophysiology setting. The association of these genotypes with risk was also confirmed when the studies were stratified according to geographic localization. The global meta-analysis showed an association in all models, indicating the association of genotypes including the allele C, with the risk of developing different diseases. When comparing between subdivisions, the CC genotype had a stronger effect on the risk to develop diseases from the autoimmune and infectious subdivisions than those in the age-related and autoimmune-inflammatory subdivisions. Moreover, the allelic model did not provide significant information on this polymorphism, indicating that the influence on determined zones or diseases is not dependent on the presence of the C allele, but rather on the specific genotype. Limitations in the present study should be noted, as in any meta-analysis, being the principal limitation the low number of studies in certain subdivisions, such as African and North American. Another limitation to consider is that the Holm-Bonferroni correction applied to validate the statistical significance of the All of the authors significantly contributed to this work. OP and MR-S designed the study; OP conducted the study selection and data extraction; OP, JG-V, LG-V, and TG performed the data analysis, genetic models assessment and prepared the figures and tables; OP and MR-S drafted the manuscript; all authors revised it critically for important intellectual content and approved the final draft.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Studies of the association between angiotensin II receptor type 1 A1166C (AGTR1 A1166C) polymorphism and chronic kidney disease (CKD) risk have yielded conflicting results. We conducted a combined case-control study and meta-analysis to better define this association. The case-control study included 634 end-stage renal disease (ESRD) patients and 739 healthy controls. AGTR1 A1166C genotype was determined using polymerase chain reaction and iPLEX Gold SNP genotyping methods. The meta-analysis included 24 studies found in the PubMed and Cochrane Library databases. Together, the case-control study and meta-analysis included 36 populations . We found no association between the C allele and ESRD . Co-dominant, dominant, and recessive model results were also not significant. No known environmental factors moderated the effect of AGTR1 A1166C on CKD in our gene-environment interaction analysis. Sensitivity analysis showed an AGTR1 A1166C-CKD association in Indian populations , but not in East Asian or Caucasian populations. Additional South Asian studies will be required to confirm the potential role of this polymorphism in CKD. Chronic kidney disease (CKD) is highly prevalent worldwide –4, and iThe renin-angiotensin system (RAS) regulates blood pressure and electrolyte balance, and an over-active RAS leads to CKD . RAS funPrevious studies found associations between AGTR1 A1166C and coronary artery disease  and breaCase and control patient characteristics are shown in Table A control sample cannot be genotyped, so the genotyping call rate in this study is 99.9%. Now there are only 634 cases and 738 controls in further analysis. We used four types of genetic models to test the association between AGTR1 A1166C and ESRD; the result is shown in Table et al. [et al. [et al. [et al. [The study identification process is shown in Figure et al. , Kim, et [et al. , Mollste [et al. , and Sha [et al.  reportedet al. [An AGTR1 A1166C and CKD forest plot was calculated using the allele model Figure . The oveet al.  reportedet al. . Additioet al. . Thus, wet al. , 34 fromWe used meta-regression to detect potential gene-environment interactions that could explain the high heterogeneity among these studies Table . None ofBoth our case-control study and meta-analysis revealed no association between AGTR1 A1166C and ESRD. Subgroup analyses identified an association between AGTR1 A1166C and CKD among Asian populations, but this was not significant in East Asians. Although patients with the AGTR1 A1166C A allele exhibit higher miR-155 and AGTR1 levels than those with the C allele , and thiThe AGTR1 A1166C-CKD association in the Asian patient subgroup might be attributable to the Indian population, based on sensitivity analysis. Although previous meta-analyses included Indian studies within the Asian subgroup , 24, theThis study had several limitations. First, we relied on tabular data for the meta-analysis, rather than individual patient data. However, a previous study revealed that the inclusion of summary data could increase the sample size and improve confidence . AdditioIn conclusion, our meta-analysis found that the AGTR1 A1166C-CKD association was only significant in the Indian subgroup, and more South Asian studies are needed to confirm this finding. The AGTR1 A1166C C allele is likely not a CKD risk factor in East Asians and Caucasians. The observed high level of heterogeneity among studies suggests a complex underlying mechanism that appears unrelated to patient sex, age, BMI, DM, hypertension, or smoking habits. Because this unexplainable heterogeneity might be related to gene-gene interactions, an AGTR1 A1166C-CKD epistasis analysis should be conducted. Some RAS polymorphisms, such as angiotensin converting enzyme I/D and angiotensinogen M235T, may moderate the association between AGTR1 A1166C and CKD. Finally, potential gene-environment interactions may influence this association, and additional environmental factors should be investigated.Based on the following thresholds, the minimum required study sample size was 1047 subjects: two-sided test with a power (1−β)=0.8 at a significance level of 0.05, ratio of controls to cases =1, hypothetical proportion of controls with exposure =9, and least extreme odds ratio (OR) to be detected =1.5 . We init2 as calculated using the MDRD equation; (2) no symptoms of kidney damage ; (3) no serious diseases (e.g. cancer), (4) able to provide a sufficient blood sample for genotyping. Patient demographic data, including age, sex, body mass index (BMI: kg/m2), history of hypertension, history of diabetes mellitus (DM), and smoking habits, were collected from medical records. Biochemistry laboratory values, such as total cholesterol, triglyceride, and creatinine levels, were collected from electronic health records. In total, 634 cases (296 men and 338 women) and 739 controls (298 men and 441 women) participated in this study up to July 2015.The case group was recruited from TSGH dialysis centers. All included patients were undergoing dialysis treatment and had been diagnosed with end-stage renal disease (ESRD). We recruited controls at the TSGH Health Management Center from patients participating in a check-up program beginning in March 2011. Control inclusion criteria were as follows: (1) estimated glomerular filtration rate (eGFR) of >60 ml/min/1.73 mGenomic DNA was extracted from peripheral blood samples using standard proteinase K  digestion and phenol/chloroform extraction methods. AGTR1 A1166C polymorphisms were genotyped using the iPLEX Gold SNP genotyping method . To valit test and reported as means ± standard deviations (SDs). Genotypes and allelic frequencies were compared between dialysis patients and healthy controls using the χ2 test or Fisher's exact test as appropriate. Logistic regression was used to estimate ORs and 95% confidence intervals (CIs) as measures of the association with CKD risk. Allele type, co-dominant, and dominant/recessive models were used to calculate the association between genetic polymorphism and CKD risk. P<0.05 was considered significant. Statistical analyses were conducted using R software, version 3.3.1 .Continuous variables were evaluated using Student's The PRISMA checklist and Meta-analysis of Genetic Association Studies Checklist are described in All studies that assessed the association between AGTR1 A1166C and CKD risk were considered for inclusion using the following criteria: (1) cross-sectional surveys or case-control studies; (2) study population aged >18 years; (3) CKD defined according to the National Kidney Foundation as kidney damage by clinical diagnosis or a glomerular filtration rate <60 ml/min/1.73 m2; (4) inclusion of at least one control group with normal kidney function; and (5) detailed genotype information. Common genetic association study methods included co-dominant, allele, dominant, and recessive models, although some papers only presented significant results. We only included papers with detailed information about genotypes. Studies that investigated relationships between genetic polymorphisms and other kidney diseases  were excluded. For manuscripts with incomplete published data, we attempted to contact the authors for further information.We recorded the first author's name, publication year, study population ethnicity, kidney functions of cases, definition of the case group, and case group population characteristics . DM and hypertension were defined as plasma glucose >126 mg/dL and systolic blood pressure >140 mmHg, respectively. If the article did not report DM and hypertension prevalence, or the definitions did not match, we assumed normal plasma glucose level and systolic blood pressure distributions for calculation purposes.Bias risk was assessed via the Cochrane Collaboration, as suggested by the Newcastle–Ottawa Quality Assessment Scale . This to2 statistic, estimated according to the DerSimonian-Laird method, was used to assess heterogeneity, and a random-effects model was used to calculate weighed effect size. Three common genetic models  were used to calculate the association between AGTR1 A1166C and CKD risk.Population characteristics of each included study are presented as means or proportions as appropriate. Our meta-analysis examined the association between AGTR1 A1166C and CKD risk in each study using ORs with 95% CIs. The τ2 value was calculated using the Cochrane Q test and used to quantify heterogeneity; an I2 value >50% indicated moderate to high heterogeneity. A meta-regression analysis of average summary values was used to explore the source of heterogeneity. According to our previous studies, the average summary value of a case group can be used to build a model and can facilitate interaction effect estimation [Egger's regression and a funnel plot were used to test the symmetry of pooled results. The Itimation . An inteP<0.05 was considered significant. However, because of multiple comparison correction, P<0.05/10 was considered significant for meta-regression analyses. Statistical analyses were conducted using the “metafor” and “meta” R version 3.3.1 packages.


Background and Objective: Studies suggests that matrix metalloproteinase (MMP)-2-1306 C/T and MMP-1-1607 1G/2G polymorphisms affect the risk of prostate cancer. However, the conclusions remain controversial and no pooled evidence of this topic has been published. Therefore, we aimed to perform a meta-analysis to shed some light on the controversial conclusion pertaining to the associations of MMP-2-1306 C/T and MMP-1-1607 1G/2G polymorphisms with prostate cancer susceptibility.Methods: A thorough literature search was performed up to August, 2016 with the PubMed, EMBASE, CBM, CNKI, and Wanfang databases. Odds ratios (ORs) and corresponding 95% confidence intervals (95% CIs) were calculated to address the correlations between these polymorphisms and risk of prostate cancer.Results: The meta-analysis included six studies  on MMP-2-1306 C/T polymorphism and three studies on MMP-1-1607 1G/2G polymorphism (438 patients and 394 controls), respectively. The overall results of meta-analysis showed that an elevated risk of the disease was implicated in MMP-2-1306 C/T polymorphism under two genetic models  and no significant association was observed between MMP-1-1607 1G/2G polymorphism and the risk of prostate cancer. The subgroup analysis results of MMP-2-1306 C/T polymorphism were similar to the overall results. However, decreased risk of prostate cancer was observed in the Caucasians for MMP-1-1607 1G/2G polymorphism.Conclusions: Current meta-analysis indicates that MMP-2-1306 C/T polymorphism is associated with elevated risk of prostate cancer, but MMP-1-1607 1G/2G polymorphism may inhibit the occurrence of prostate cancer in Caucasians. Further studies are warranted to verify the conclusions. Prostate cancer (MIM 176807) is the most common malignancy in males diagnosed in western countries, which is also the second leading cause of death from cancer among men in the US  and MMP-1-1607 1G/2G (rs1799750) polymorphisms and risk of prostate cancer  are a family of proteolytic enzymes that degrade extracellular matrix proteins. MMPs degrade basement membranes and extracellular matrix; the processes are essential for angiogenesis and invasion of carcinoma. More than 20 members in the MMP family have been found in humans  statement in reporting meta-analysis  and risk of prostate cancer; (5) study subjects were human beings. The exclusion criteria were as following: (1) review articles, meta-analysis or systematic review, and case reports; (2) the genotype distribution was not available; (3) duplicated publications (studies recently published or with more subjects were included).Studies were included if they met the following criteria: (1) the case-control study design was performed; (2) patients were diagnosed with prostate cancer and controls were cancer-free; (3) genotype distribution was presented to calculate odds ratios (ORs) with corresponding 95% confidence intervals (95% CIs); (4) study investigated the association between Titles and abstracts of papers yielded from online databases based on the prespecified search strategy were screened independently by two reviewers (HW and XTZ) to identify studies that met the prespecified eligibility criteria. The same two reviewers (HW and XTZ) independently retrieved and assessed the potentially eligible publications. Any discrepancy between them over the eligibility of particular articles was settled in consultation with a third reviewer (XHW).A standardized, pre-specified form was used to extract data from the included papers for pooled analyses. Extracted information was as following: surname of the first author, publication year, region of the study, ethnicity of the participants, sample size, source of control, genotyping method, distribution of gene frequency, minor allele frequency (MAF), allelic gene frequency, Hardy-Weinberg equilibrium (HWE). Two review authors extracted data independently, and discrepancies were resolved through discussing with a third author if necessary. Missing data would be requested from study authors through e-mail address presented in the paper.2 test and the I2 statistic. The I2 value greater than 50% and/or P-value less than 0.1 were indicative of substantial heterogeneity. Sensitivity analysis was conducted based on HWE  of MMP-1-1607 1G/2G polymorphism, so fixed-effects model was applied for meta-analysis. Nevertheless, moderate to high between-study heterogeneity was detected in the remaining four genetic models of MMP-1-1607 1G/2G polymorphism  and TT+CT vs. CC model . No significant association was observed under the contrasts of T vs. C . Subgroup analyses showed that an elevated risk of prostate cancer was observed in Caucasians .The results of meta-analysis involving MMP-1-1607 1G/2G polymorphism decreased the risk of prostate cancer among Caucasians under 2G vs. 1G model  and 2G/2G vs. 1G/1G+1G/2G model .The results of sensitivity analysis were similar to the overall analyses by excluding one study in which the controls were not consistent with HWE. Subgroup analysis based on ethnicity showed that MMP-2-1306 C/T polymorphism. The Egger's publication bias plot . Due to limited quantity of included studies on MMP-1-1607 1G/2G polymorphism, we did not test the publication bias for this polymorphism.The publication bias was detected using Egger's line regression test for T vs. C model of MMP-2-1306 C/T and MMP-1-1607 1G/2G polymorphisms and risk of prostate cancer. In this meta-analysis, we identified six studies for MMP-2-1306 C/T polymorphism and three studies for MMP-1-1607 1G/2G polymorphism. The overall analyses showed that the CT and TT+CT carriers presented 1.78- and 1.62-fold higher risk of prostate cancer than CC carriers, respectively. Sensitivity and subgroup analyses showed similar results to the overall analyses, except allelic genetic model. After excluding studies in which controls were not consistent with HWE, the result tended to be significant in Caucasians under the allelic model. The overall analyses of MMP-1-1607 1G/2G polymorphism and risk of prostate cancer showed no statistical significance. However, decreased risk of prostate cancer was observed in the Caucasians for MMP-1-1607 1G/2G polymorphism.The present meta-analysis, for the first time, comprehensively evaluated the associations between MMP-1-1607 1G/2G polymorphism in Caucasians might be spurious results or false positives. The significant differences were observed based on only two studies with limited participants. Therefore, the statistical power was relatively weak. Additionally, mutation is usually to be harmful for human and little variation tends to be beneficial. Hence the results indicated that more studies with large sample size were necessary to be carried out for investigating the association.The decreased risk of prostate cancer concerning MMP-2-1306 C/T and MMP-1-1607 1G/2G polymorphisms with the risk of prostate cancer using meta-analysis. Previous studies were all original case-control studies, which had limited statistical power. Therefore, the results of our study were more precise than that of previous studies. In addition, there are many other polymorphisms in the MMP-2, MMP-1 and other MMPs genes. Therefore, future studies should take these polymorphisms into consideration and haplotype analyses are required to be performed to give more precise results. The prostate cancer is mainly diagnosed by PSA test in clinical practice, and this test has become increasingly prevalent during the last decade. Therefore, the majority of prostate cancer cases are diagnosed with organ-confined cancer (Wirth et al., MMP-2-1306 C/T polymorphism and the risk of prostate cancer was truly effective, researchers ought to explore the association between the polymorphism and the risk of aggressive prostate cancer, which may be a new predicting biomarker for the life-threatening disease.The most important advantage of our study is that this is the first study to examine the associationsofMMP-1-1607 1G/2G polymorphism and risk of prostate cancer. Therefore the statistical power was relatively limited (Zeng et al., MMP-2 and MMP-1 genes and other risk genetic polymorphisms might also impact the level of MMPs, such as rs1477017 and 17301608 in MMP-2 gene (Jacobs et al., MMP-2-1306 C/T polymorphism, the publication bias could not be avoided. Furthermore, the most populations included in the meta-analysis were in Caucasian populations, therefore the external validity is limited. Lastly, moderate to high between-study heterogeneity was detected for MMP-1-1607 1G/2G polymorphism, which might distort the results. Therefore, the results of association between MMP-1-1607 1G/2G polymorphism and risk of prostate cancer should be interpreted with caution.Several limitations should also be taken into consideration when interpreting the results of this meta-analysis. First, the quantity and sample size of included studies were relatively small even though we undertook a comprehensive literature search, especially for the association between MMP-2-1306 C/T polymorphism is associated with elevated risk of prostate cancer, but MMP-1-1607 1G/2G polymorphism may have an inhibitory effect on the risk of prostate cancer in Caucasians. More studies with large sample size are warranted to further verify the conclusions in future.In summary, our findings suggest that the HW and X-HW: Conceived and designed the study; HW, X-TZ, and X-HW: Participated in study selection, data extraction; HW, T-ZL, and D-LH: Performed statistical analysis; HW, X-TZ, and X-HW: Were involved in manuscript drafting and revision. All authors approved the final manuscript for submission and publication.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Objective To evaluate the strength and validity of the evidence for the association between adiposity and risk of developing or dying from cancer.Design Umbrella review of systematic reviews and meta-analyses.Data sources PubMed, Embase, Cochrane Database of Systematic Reviews, and manual screening of retrieved references.Eligibility criteria Systematic reviews or meta-analyses of observational studies that evaluated the association between indices of adiposity and risk of developing or dying from cancer.Data synthesis Primary analysis focused on cohort studies exploring associations for continuous measures of adiposity. The evidence was graded into strong, highly suggestive, suggestive, or weak after applying criteria that included the statistical significance of the random effects summary estimate and of the largest study in a meta-analysis, the number of cancer cases, heterogeneity between studies, 95% prediction intervals, small study effects, excess significance bias, and sensitivity analysis with credibility ceilings.Results 204 meta-analyses investigated associations between seven indices of adiposity and developing or dying from 36 primary cancers and their subtypes. Of the 95 meta-analyses that included cohort studies and used a continuous scale to measure adiposity, only 12 (13%) associations for nine cancers were supported by strong evidence. An increase in body mass index was associated with a higher risk of developing oesophageal adenocarcinoma; colon and rectal cancer in men; biliary tract system and pancreatic cancer; endometrial cancer in premenopausal women; kidney cancer; and multiple myeloma. Weight gain and waist to hip circumference ratio were associated with higher risks of postmenopausal breast cancer in women who have never used hormone replacement therapy and endometrial cancer, respectively. The increase in the risk of developing cancer for every 5 kg/m2 increase in body mass index ranged from 9%  for rectal cancer among men to 56%  for biliary tract system cancer. The risk of postmenopausal breast cancer among women who have never used HRT increased by 11% for each 5 kg of weight gain in adulthood , and the risk of endometrial cancer increased by 21% for each 0.1 increase in waist to hip ratio . Five additional associations were supported by strong evidence when categorical measures of adiposity were included: weight gain with colorectal cancer; body mass index with gallbladder, gastric cardia, and ovarian cancer; and multiple myeloma mortality.Conclusions Although the association of adiposity with cancer risk has been extensively studied, associations for only 11 cancers  were supported by strong evidence. Other associations could be genuine, but substantial uncertainty remains. Obesity is becoming one of the biggest problems in public health; evidence on the strength of the associated risks may allow finer selection of those at higher risk of cancer, who could be targeted for personalised prevention strategies. Cancer is a leading cause of death worldwide, with an estimated 12.7 million new cases and 7.6 million deaths from cancer annually.Several meta-analyses support the link between obesity and cancer, but substantial heterogeneity exists between studies.To summarise and evaluate the existing evidence and appraise its quality, we performed an umbrella review of systematic reviews and meta-analyses that investigated the association between adiposity indices and risk of developing or dying from cancer. Umbrella reviews systematically appraise the evidence on an entire topic across many meta-analyses of multiple putative risk factors on multiple outcomes.We searched PubMed, Embase, and the Cochrane Database of Systematic Reviews from inception to May 2015 for systematic reviews or meta-analyses that investigated the association between adiposity indices and risk of developing or dying from any cancer using a predefined search algorithm . Adiposity indices were body mass index, waist circumference, hip circumference, waist to hip ratio, weight, weight gain, and weight loss from bariatric surgery. We hand searched the references of the retrieved systematic reviews and meta-analyses and the proceedings of relevant conferences for articles missed by the electronic search and unpublished data.We included systematic reviews and meta-analyses of observational epidemiological studies in humans that assessed continuous or categorical measures of adiposity. We excluded studies in which physical activity, height, or prenatal factors  were the exposures of interest and those in which cancer incidence or mortality were not the outcomes of interest. We excluded meta-analyses of prognostic studies that investigated cancer survival or other outcomes among patients with cancer and meta-analyses that did not include comprehensive data from individual studies—such as relative risks, 95% confidence intervals, number of cases and controls, or total population. When we found more than one meta-analysis on the same association between exposure and outcome, we included the one with the largest number of primary studies to avoid duplication. We subsequently examined whether the summary estimates showed concordance between the included and duplicate meta-analyses.We extracted the name of the first author, year of publication, outcome, and exposure from each eligible systematic review or meta-analysis. From each individual study in a meta-analysis, we extracted the first author, year of publication, epidemiological design, number of cases and controls in case-control studies or total population in cohort studies, maximally adjusted relative risk  and 95% confidence intervals. Two investigators  independently searched the literature, assessed the eligibility of the retrieved papers, and extracted the data. Disagreements were resolved by discussion with a third investigator (KT).Estimation of summary effect—For each exposure and outcome pair, we calculated the summary effect and the 95% confidence interval using fixed and random effects methods.Assessment of heterogeneity—The heterogeneity between studies was assessed with Cochran’s Q test2 statistic.Estimation of prediction intervals—To further account for heterogeneity between studies we calculated 95% prediction intervals for the summary random effect estimates, which represent the range in which the effect estimates of future studies will lie.Assessment of small study effects—We examined whether smaller studies gave higher risk estimates than larger studies, which is an indication of publication bias, true heterogeneity, or chance.Evaluation of excess significance—We assessed excess significance bias by evaluating whether the observed number of studies with nominally statistically significant results  in the published literature was different from the expected number of studies with statistically significant results.t distribution.Credibility ceilings—We used credibility ceilings, a sensitivity analysis tool, to account for potential methodological limitations of observational studies that might lead to spurious precision of combined effect estimates.c (credibility ceiling) that the true effect size is in a different direction from the one suggested by the point estimate. The pooled effect size and the heterogeneity between studies were re-estimated using a wide range of credibility ceiling values.−6, a threshold that might substantially reduce false positive findings2 statistic of heterogeneity between studies is smaller than 50%, the 95% prediction intervals excludes the null value, small study effects or excess significance bias are not indicated, and the association survives a credibility ceiling of at least 10%.The associations between measures of adiposity and cancer were categorised into strong, highly suggestive, suggestive, or weak depending on the strength and validity of the evidence.−6, the meta-analysis had more than 1000 cases, and the largest study in the meta-analysis was nominally statistically significant. The criteria for suggestive were: random effects P<10−3 and the meta-analysis had more than 1000 cases. All other associations with a nominally statistically significant P value for the random effects meta-analysis were deemed to provide weak evidence.The association was considered highly suggestive if P<10The primary analysis in this umbrella review focused on cohort studies using continuous measures of adiposity, which was considered a more valid and standardised way of presenting and synthesising individual study estimates than using categorical measures. Sensitivity analyses were conducted using case-control studies and categorical measures. All statistical analyses were performed using Stata version 13 ,No patients were involved in setting the research question or outcome measures, nor were they involved in developing plans for design or implementation of the study. No patients were asked to advise on interpretation or writing up on results. The results will be disseminated to the general public through public presentations and the authors’ involvement in different charities.Overall, 49 eligible papers were identified that included a total of 204 meta-analyses. These meta-analyses summarised 2179 individual study estimates from 507 unique cohort or case-control studies . These 95 meta-analyses studied 28 different cancer sites or subtypes and included 818 individual study estimates, with body mass index in adults being investigated most frequently .The 204 meta-analyses included associations between seven indices of adiposity  and the incidence (n=196) or mortality (n=8) from cancer at 36 anatomical sites and by subtypes. A total of 194 meta-analyses included cohort studies (see supplementary tables 1 and 2). Of these, 95 used a continuous scale to measure adiposity .−6, 54 (57%) and 35 (37%) meta-analyses were significant, respectively.When we used a threshold of P<0.05, the summary fixed effects estimates were significant in 76 of the 95 meta-analyses (80%) and the summary random effects were significant in 72 (76%) (see supplementary table 1 and supplementary figures 1-28). At a stricter threshold of P<0.001, 66 (69%) and 59 (62%) meta-analyses produced significant summary results using the fixed and random effects models, respectively. At P<10Of those 35 meta-analyses, 31 found an increased risk of cancer with higher adiposity for oesophageal adenocarcinoma, multiple myeloma, and cancers of the colon, rectum, liver, biliary tract system , pancreas, postmenopausal breast, endometrium, and kidney. Four of the 35 meta-analyses indicated an inverse association for oesophageal squamous cell carcinoma and lung cancer. The magnitude of the observed summary random effect estimates ranged from 0.57 to 1.90; 61% of the estimates lay between 0.80 and 1.20  and 1.09 (1.06 to 1.13), respectively) than in women (1.12 (1.06 to 1.17) and 1.02 (0.99 to 1.05), respectively). This was also true for other adiposity indices. Moreover, body mass index was associated with an increased risk of melanoma in men  but not in women .The summary estimates were similar between men and women for eight of 11 cancers . However, the incidence of colon and rectal cancer was higher per 5 kg/m2=50-75%) and 15 meta-analyses (16%) showed substantial heterogeneity (I2 >75%) for some adiposity indices and several cancers . We further assessed the uncertainty of the summary effects by calculating 95% prediction intervals; the null value was excluded in only 27 associations.Τhe Q test was significant at P≤0.10 in 42 of 95 meta-analyses (44%) (see supplementary table 2). Twenty meta-analyses (21%) showed considerable heterogeneity . Supplementary figures 1 to 28 show the funnel plots for each cancer site.Nineteen meta-analyses had evidence of excess significance bias for some adiposity indices using the largest study estimate as the plausible effect size . Using summary fixed or random effects estimates as plausible effect sizes gave similar results.2 estimate larger than 50% (supplementary table 4).A credibility ceiling of 0% corresponds to the random effects model calculation. Of the 95 meta-analyses, 72 (76%) retained nominal statistical significance (P<0.05) with a credibility ceiling of 5%. Fifty (53%), 33 (35%), and 19 (20%) meta-analyses remained statistically significant with ceilings of 10%, 15%, and 20%, respectively. Heterogeneity between studies gradually decreased with increasing ceilings. With a ceiling of 10%, no meta-analyses of cohort studies with continuous exposure gave an IWe explored whether the reported associations between adiposity and developing or dying from cancer were supported by strong, highly suggestive, suggestive, or weak evidence  for colorectal cancer among men to 56%  for biliary tract system cancer. The risk of postmenopausal breast cancer among women who have never used HRT increased by 11% for each 5 kg of weight gain , and the risk of endometrial cancer increased by 21% for each 0.1 increase in waist to hip ratio .Only 13% of meta-analyses (12 of 95) were supported by strong evidence. They summarised data on body mass index (n=10), waist to hip ratio (n=1), and weight gain (n=1). An increase in body mass index was associated with a higher risk of developing oesophageal adenocarcinoma; colon and rectal cancer in men; biliary tract system and pancreatic cancer; endometrial cancer in premenopausal women; kidney cancer; and multiple myeloma. Weight gain and waist to hip circumference ratio were associated with higher risks of postmenopausal breast cancer in women who have never used hormone replacement therapy and endometrial cancer, respectively. The increase in the risk of developing cancer for every 5 kg/mSeventeen meta-analyses (18%) were supported by highly suggestive evidence, and they found positive associations for colon (with body mass index and waist circumference), liver (with body mass index), postmenopausal breast (with body mass index), total endometrial , postmenopausal, type I and type II endometrial (with body mass index), and kidney cancer (with body mass index). Inverse associations were found for body mass index in adults with oesophageal squamous cell carcinoma  and lung cancer . Twenty four meta-analyses (25%) were supported by suggestive evidence for an association, and 19 meta-analyses (20%) were supported by weak evidence.The evidence was graded similarly in the 194 meta-analyses of cohort studies that assessed continuous and categorical measures of adiposity (supplementary table 6). Five additional associations were supported by strong evidence: weight gain with incidence of colorectal cancer, body mass index with incidence of gallbladder, gastric cardia, and ovarian cancers, and mortality from multiple myeloma. When both cohort and case-control studies were assessed for continuous and categorical measures of adiposity (supplementary table 7), two additional associations were supported by strong evidence: body mass index and risk of melanoma and meningioma, both of which were judged as weak associations when only cohort studies were evaluated. The strong evidence for an association between body mass index and incidence of biliary tract system cancerwas downgraded to suggestive evidence when both case-control and cohort studies were evaluated.We reviewed 204 meta-analyses to evaluate the current evidence for associations between seven adiposity indices and the risk of developing or dying from 36 primary cancers and their subtypes. Twelve associations, stemming from cohort studies using continuous measures of adiposity, were supported by strong evidence, inferred by strongly statistically significant results and no suggestion of bias. These associations were primarily between body mass index and malignancies of digestive organs , biliary tract system, and pancreas), hormone related cancers , premenopausal and overall endometrial cancer, kidney cancer, and multiple myeloma. Five additional associations were supported by strong evidence when categorical measures of adiposity were used: weight gain with risk of colorectal cancer risk and body mass index with risk of gallbladder, gastric cardia, and ovarian cancer, and mortality from multiple myeloma.−6) were used, the proportion of significant associations decreased to 37%. Large heterogeneity (I2 ≥50%) was observed in 37% of the meta-analyses. When we calculated the 95% prediction intervals, which further account for heterogeneity, we found that the null value was excluded in only about one third of the associations. Moreover, some meta-analyses had evidence of small study effects or excess significance bias. Most meta-analyses (53%) preserved their statistical significance with a 10% credibility ceiling, but only one in five preserved significance at a ceiling of 20%, indicating that many associations between adiposity and cancer remain uncertain.The effect of obesity on the incidence and mortality of cancer is well recognisedOur grading of the evidence largely agrees with systematic analyses of the literature performed by the World Cancer Research Fund (WCRF) and the International Agency for Research on Cancer (IARC).Several methods exist for rating evidence, but they are inconsistent and allow some degree of arbitrariness.Most of the associations between adiposity indices and endometrial cancer were supported by strong or highly suggestive evidence. In particular, the association between waist to hip ratio and risk of total endometrial cancer was supported by strong evidence, indicating that central obesity, which is linked to hyperinsulinaemia and type 2 diabetes, has a major role in the development of this disease. Strong evidence also supported the association between body mass index and premenopausal endometrial cancer. We found the associations between body mass index in adults or at around age 20, waist circumference, weight, and weight gain and total endometrial cancer, and between body mass index and postmenopausal type I and II endometrial cancer to be highly suggestive owing to substantial heterogeneity between studies and potential for excess significance bias. These results were in agreement with the WCRF and IARC, which did not provide separate ratings for menopausal status or histological subtype of disease.We found strong evidence to support the association between weight gain in adulthood and postmenopausal breast cancer among women who have never used HRT, yielding an 11% higher risk per 5 kg of weight gain. We found highly suggestive evidence to support the positive association of body mass index with postmenopausal breast cancer, irrespective of HRT. Weight gain may be a better metric than body mass index for measuring the dynamic nature of adiposity during adulthood, when obesity becomes central and has more metabolic effects.2 increase in body mass index. The risk was 12% higher for each 5 kg/m2 increase in body mass index among women in a meta-analysis of 20 cohorts that showed substantial heterogeneity and evidence of small study effects and excess significance bias. The association between body mass index and rectal cancer in men was supported by strong evidence, but with a summary relative risk considerably smaller than for colon cancer (1.09 v 1.30). We found no association between body mass index and rectal cancer in women. The evidence for most of the other adiposity indices was suggestive, as few studies have investigated these associations. The insulin signalling pathway is a possible mechanism underlying the association between obesity and colorectal cancer in men.The association between body mass index and colon cancer was supported by strong evidence in men and suggestive evidence in women. Twenty four cohorts investigated the association in men, totalling a 30% higher risk of colon cancer for each 5 kg/mWe found that higher body mass index was associated with an increased risk of oesophageal adenocarcinoma, and this was supported by strong evidence. We found an inverse association for oesophageal squamous cell carcinoma, which was supported by highly suggestive evidence due to considerable heterogeneity between studies. Similar summary associations were found by the WCRF and IARC, which concluded that adiposity convincingly increased the risk of oesophageal adenocarcinoma.We found an inverse association between body mass index and lung cancer that was supported by highly suggestive evidence but had substantial heterogeneity between studies and evidence of small study effects. When we investigated this association by smoking status, we found an inverse association between body mass index and lung cancer among smokers, which was supported by suggestive evidence. We found a non-significant association among non-smokers. These results may be due to residual confounding, reverse causation, or effect modification by smoking, which led the WCRF and IARC to grade this evidence as inadequate.Furthermore, we found that the associations between body mass index and multiple myeloma and cancers of the biliary tract system, pancreas, and kidney were supported by strong evidence. The WCRF and IARC ratings were similar, except the WCRF did not study multiple myeloma.Our review relied on previously published meta-analyses and the literature searches performed by the authors of those meta-analyses. Some studies may have been missed, although this is unlikely to have influenced our findings because our assessment of duplicate meta-analyses on the same exposure-outcome associations gave similar summary results. Some of the meta-analyses that used continuous measures of adiposity were published as long ago as 2008, which may mean that we didn’t include the most recent evidence. However, we found that the evidence grading for recently published meta-analyses investigating the association of categorical terms of body mass index with the same cancers was similar.We evaluated all study specific results that were reported in the meta-analyses , but we may have missed other subanalyses that were not reported with sufficient study specific data. Assessing the quality of the primary studies included in the meta-analyses was beyond the scope of this umbrella review. Finally, the statistical tests we used to assess bias do not prove its definitive presence or its exact source. However, our estimates are likely to be conservative, as a negative test does not exclude the potential for bias. Moreover, our evidence grading was not sensitive to the use of 95% prediction intervals, excess significance tests, or credibility ceilings, because when we consecutively removed these criteria from the quality of evidence algorithm, the resulting inference remained the same.The association between obesity and risk of developing or dying from cancer has been extensively studied. We found strong evidence to support the positive association between obesity and 11 of the 36 cancer sites and subtypes that we examined, predominantly comprising cancers of the digestive organs and hormone related malignancies in women. Substantial uncertainty remains for the other cancers. To draw firmer conclusions we need prospective studies and large consortiums with better assessment of the changing nature of body fatness and with comprehensive standardised reporting of analyses. As obesity becomes one of the greatest public health problems worldwide, evidence of the strength of the associations between obesity and cancer may allow finer selection of people at high risk, who could be selected for personalised primary and secondary prevention strategies.Cancer is a leading cause of death worldwide, and the prevalence of obesity has more than doubled over the past 40 yearsNumerous meta-analyses support the link between excess body weight and increased risk of developing and dying from several types of cancerThe reported associations may be causal for some malignancies, but they may be flawed owing to inherent biases that exaggerate the effect of obesity on cancer incidence and mortalityStrong evidence supported the association between obesity and only 11 of 36 cancer sites and subtypes, predominantly comprising cancers of digestive organs and hormone related malignancies in womenOther associations could be genuine, but substantial uncertainty remains


P = 0.009; GG versus AA: OR = 0.233, P = 0.002; GA versus AA: OR =0.294, P = 0.006; GG versus AA+AG: OR =0.635, P = 0; GA+GG versus AA: OR = 0.242, P = 0.003), but not in general populations and Caucasians. In conclusion, LEP G2548A polymorphism has no relationship with BC susceptibility, while LEPR Q223R polymorphism could decrease BC risk in Asians, but not in overall individuals and Caucasians. More multicenter studies with larger sample sizes are required for further investigation.Many studies have been performed to investigate the correlation of leptin (LEP) and leptin receptor (LEPR) polymorphisms with breast cancer (BC) risk, however the results are inconclusive. To obtain a more precise estimation, we conducted this meta-analysis. We searched PubMed, EMBASE, and Web of Science databases to identify qualified studies. Pooled odds ratios (ORs) with 95% confidence intervals (CIs) were used to evaluate the association. Eight eligible studies  for LEP G2548A (rs7799039) polymorphism, and thirteen studies  for LEPR Q223R (rs1137101) polymorphism were included in our study. In general, no significant association between LEP G2548A polymorphism and BC susceptibility was found among five genetic models. In the stratified analysis by ethnicity and sources of controls, significant associations were still not detected in all genetic models. For LEPR Q223R polymorphism, we observed that the association was only statistically significant in Asians (G versus A: OR = 0.532,  Breast cancer is a complex and heterogeneous disease and has become a challenging health issue confronted by women worldwide , 2. PrevLeptin is the product of the ob gene that plays a pivotal role in regulating food intake, energy expenditure and neuroendocrine function; furthermore, high levels of leptin have been frequently observed in obese subjects , 7, 8. SNevertheless, the studies about rs7799039 and rs1137101 polymorphisms with the risk of BC at present were still controversial: the study by Liu et al.  discoverThe complete search process is presented in Figure Genotype, distribution and allele frequency of the two polymorphisms are shown in Table P = 0.009), GG versus AA , GA versus AA , GG versus AA+AG  and GA+GG versus AA , while no meaningful correlation was observed in Caucasians. In addition, we did not observe any association between LEPR Q223R polymorphism and BC risk among the subgroups of population-based controls or hospital-based controls   Table . BesidesWith the continuous development of society and the progress of technology, especially the improvement in the field of medicine, the knowledge about BC is constantly updating as time goes on. Recently, obesity has been considered as one of the related factors for BC risk , 4. HoweLEP G2548A polymorphism, located in the 5’ region of the LEP gene, is implicated in transcription . SeveralLeptin exerted its downstream functioning through binding to leptin receptor . For LEPSeveral potential limitations of this meta-analysis should be addressed. First, our study was conducted without adjustment for several potential confounding variables, because of lacking information including age, smoking status, drinking status, environmental factors, and other habits related to lifestyle, which are known to have significant effects on the risk of BC , 30. SecIn summary, LEP G2548A polymorphism has no relationship with BC susceptibility, while LEPR Q223R polymorphism could decrease BC risk in Asians, but not in general individuals and Caucasians. However, given the limited sample size, more multicenter studies with larger sample sizes are necessary to clarify the relationships of these polymorphisms with BC risk.We searched PubMed and EMBASE and Web of Science database up to May 1, 2017. The searching strategy was as follows: (breast cancer OR breast carcinoma) AND (polymorphism OR variant OR genotype OR SNP) AND (leptin OR LEP OR LEPR OR leptin gene receptor). The searching was conducted without the limitations of language. Besides, the references of retrieved studies were further examined to identify additional eligible publications.The studies included in the meta-analysis must meet all the following inclusion criteria: (1) clinical trials studying the association between LEP G2548A, or LEPR Q223R polymorphism and the risk of BC; (2) case-control design; (3) sufficient data for calculation of odds ratio (OR) with confidence interval (CI); (4) all the BC subjects in case groups must be pathologically confirmed. Accordingly, the following exclusion criteria were also used: (1) not case-control studies; (2) abstracts, comments, reviews, meta-analyses, and duplicated publications; (3) no detailed genotyping data; (4) the source of cases and controls, and other essential information were not provided; (5) duplicate studies.P value for Hardy-Weinberg equilibrium (HWE).The data from the included studies were independently extracted by two investigators, and discrepancies were resolved through discussion by the research team. The following information was extracted: the first author, year of publication, country of origin, ethnicity, genotyping method, sources of controls, number of cases and controls, and 2 test and Q test. For I2 test, the criteria for heterogeneity were as follows: I2 < 25%, no heterogeneity; 25%-75%, moderate heterogeneity; I2 > 75%, high heterogeneity. If the P value of the Q test was < 0.1, the random-effects model was used; if not, we would select the fixed-effects model to carry out our research. Sensitivity analysis was performed by deleting one study at a time. To evaluate the publication bias, we used Begg's test, Egger's test and funnel plot. Subgroup analysis was performed according to ethnicity and sources of controls. The P value for Hardy-Weinberg equilibrium (HWE) was calculated by the chi-square test in the control group in each study. All statistical analyses were performed using STATA version 10.0 software . All P values were two sided, and P < 0.05 was considered statistically significant.Pooled odds ratios (ORs) and 95% confidence intervals (CIs) were used to measure the association between these two polymorphisms and BC risk among five genetic models: an allele contrast genetic model, a homozygote genetic model, a heterozygote genetic model, a dominant genetic model, and a recessive genetic model. Heterogeneity among studies was evaluated by I


Then, we conducted a case-control study of Han Chinese in central China to verify the evidence-based results.  We conducted an update meta-analysis aiming to verify the association between p27-V109G polymorphism was determined by MassARRAY genotyping method.  Relevant studies were collected from diverse databases up to March 2017. In addition, a hospital-based (H-B) case-control study enrolling 90 PCa patients and 140 healthy controls was included to verify these evidence-based findings. Genetic risk was calculated by odds ratio (OR) with its corresponding 95% confidence interval (CI). The p27-V109G polymorphism decreased overall cancer risk in allelic contrast, heterozygote, and dominant models. When stratified analysis was conducted by ethnicity, data revealed that p27-V109G polymorphism was associated with a decreased cancer risk in Caucasians. Highlighted in the subgroup analysis by cancer type, we uncovered a significantly decreased risk of PCa in allelic contrast, dominant, homogeneous, and recessive models. However, in the validation case-control study, we failed to uncover a positive association between p27-V109G polymorphism and PCa risk. In addition, negative results were also identified when subgroup analyses were stratified by age, tumor grade, tumor stage, PSA levels, and other measurements.  Finally, twenty-four published studies comprising 9627 cases and 12,102 controls were enrolled for the current meta-analysis. Overall analysis suggested that p27-V109G polymorphism plays a protective role in overall cancer risk, particularly for PCa, our case-control study failed to validate any association between this particular polymorphism and PCa risk. Although evidence-based results suggest that  In the Asian population, although PCa incidence remains at a relatively low level, it keeps increasing , 2. Therp27  controls the cell cycle check from G1 to S transition , and PSA levels (PSA ≤ 20 ng/ml and PSA > 20 ng/ml) were subcategorized into 2 subgroups. However, no significant association was also observed in all these subgroups .As we all know, etiology of PCa remains unclear. It has been postulated that many factors can result in malignant neoplasms. In recent decades, many researchers drew attention to the association between genetic polymorphisms and cancer susceptibility , 43. Sinp27 play an inherited role in cancer. p27 mainly include two polymorphisms [p27 polymorphism has a protective effect on cancers [p27 rs34330 polymorphism and cancer risk; they suggested that rs34330 polymorphism was related to an increased risk of cancer. Here, we reconducted an updated meta-analysis containing 24 case-control studies comprising of 9627 cases and 12,102 controls and identified that p27-V109G polymorphism was associated with a decreased risk of cancer, which was consistent with Lu et al.'s work [p27-V109G polymorphism and cancer risk in Asians, but it seemed to work as a protective effect in Caucasians. Moreover, in stratification analysis by cancer type, we also uncovered a significant decreased risk of PCa.According to the literatures, several variants located in orphisms , called  cancers , 32, 33, cancers , 46, 47. cancers  conducte cancers  also con.'s work . In addip27-V109G polymorphism played a protective role in PCa risk. However, results from our case-control study were not consistent with the studies mentioned above, as well as the evidence-based results derived from meta-analysis, but consistent with Huang et al.'s [p27-V109G polymorphism was not associated with PCa. In addition, our results also suggested that no significant associations were also uncovered between this polymorphism and clinicopathologic features of PCa. Hereditary heterogeneity of subjects, different polymorphisms of p27, or inadequate adjustment for confounding factors can lead to the above inconsistency. Apart from these, sample size may also affect the results.Previously, several publications have studied the association between rs2066827 polymorphism and PCa risk in different ethnicities , 50, 51.p27-V109G polymorphism and PCa risk, the enrolled sample size was relatively small, which may contribute to false-positive or false-negative findings. Thirdly, we only took the p27 genetic variant into account; other genetic factors, such as gene-gene and gene-environment interactions, can also influence cancer risk. Fourthly, for subgroup analysis, lack of adequate eligible studies may lead to insufficient statistical power, limiting the ability to identify weakly positive results. Fifthly, for the case-control study, we only made a genetic analysis of the p27-V109G in a Chinese population and the sample size was also limited. Moreover, the age of the cases and controls and the existing statistical significance may affect the results of the current work.Although we have carried out a comprehensive retrieval on diverse databases, and obtained a convincible evidence-based results, which were also validated by a H-B based case-control study, there are still several limitations that should be noted in our study. First, only studies published in English or Chinese were included, which will result in potential study selection bias. Second, for PCa, they are only three case-control studies included and, of them, two generated from an Asian (Chinese) population. Although we obtained a positive association between p27-V109G polymorphism may not be the risk factor for PCa, a result inconsistent with our meta-analysis work. Although meta-analysis is a powerful tool to cumulate and summarize the knowledge in a research field through statistical instruments, it can also be limited by the number of studies and sample sizes enrolled. Nevertheless, all these findings may require larger sample-size studies in the Chinese Han population to further validate them.Taken together, the current work suggests that


Polymorphic variants of genes involved in folate metabolism are implicated in the susceptibility to meningioma and glioma, but the results from published articles are controversial and inconclusive. Therefore, we performed this meta-analysis including all studies available to evaluate the relationship between folate metabolism genetic polymorphisms and the susceptibility to meningioma and glioma in adults. We searched the literature in PubMed, EMBASE and Cochrane Central Library for relevant articles published up to August 2016. The odds ratios (ORs) and the corresponding 95% confidence intervals (95%Cls) were used to evaluate the associations of two folate metabolism genetic variants MTRR A66G (rs1801394) and MTHFR A1298C (rs1801131) with the risk of meningioma and glioma in adults. We found significant association of MTHFR A1298C (rs1801131) variant genotypes with increased incidence of meningioma and glioma in this study population . Moreover, we found that MTRR A66G (rs1801394) variant genotypes was associated with increased risk of meningioma and glioma . In conclusion, our meta-analysis suggests that two folate metabolism genetic variants MTRR A66G (rs1801394) and MTHFR A1298C (rs1801131) contribute to genetic susceptibility to meningioma and glioma in adults. Based on the GLOBOCAN2012 investigations, approximately 14.1 million new cancer cases and 8.2 million deaths were reported worldwide . PrimaryFolate metabolism plays an important role in carcinogenesis, due to its involvement in DNA synthesis, methylation and repair. Folate metabolism regulates nucleotide synthesis and DNA methylation via a complex pathway involving at least 30 different enzymes . TherefoSemmler et al. reported the first case-control study showing that A1298C genetic variant was not significantly associated with brain tumor susceptibility . We hypoAfter screening the abstracts, titles, or contents through EMBASE, PubMed and the Cochrane Library, we identified 72 potentially relevant studies and selected five published studies –16. The Meta-analysis of A66G polymorphism in 2,236 cases and 2,248 controls showed a significant association between A66G and the risk of meningioma and glioma   ; otherwiBoth Begg’s and Egger’s tests were used to evaluate the publication bias of the studies , 20. TheAccumulating evidences indicate that the intake of folic acid has a negative correlation with cancer –23, inclHuman MTHFR gene is located on 1p36.3 locus and contains 11 exons and 10 introns, encoding 656 amino acids , 28. MTHAlthough increased number of case-control studies investigated the association of folate metabolism genetic variants with brain tumor susceptibility in adults, the results are inconclusive. Numerous studies have reported that folate metabolism genetic variants are associated with the risk of several cancers, including head and neck , lung 3, breast The findings of current study should be interpreted with caution due to several potential limitations. First, the majority of the subjects included in present study were ethnically Caucasian, thus subjects from more diverse ethnicities should be included in future studies. Second, the number of subjects enrolled in certain subgroups was relatively small. Owing to the lack of the original data, we could not estimate meningioma and glioma susceptibility stratified by the gender, age, life-style and other risk factors. Data from large-scale multi-center studies are needed to verify the association between folate metabolism genetic variants (A1298C and A66G) and meningioma and glioma susceptibility in adults.In conclusion, our findings suggest that folate metabolism genetic variants (A1298C and A66G) may increase the susceptibility of meningioma and glioma in adults. Further large-scale, multi-center and well-designed studies are necessary to investigate the potential function of genetic variation of folate metabolism in meningioma and glioma in adults.We carried out a literature search in EMBASE, PubMed, and the Cochrane Library up to August 2016. We performed electronic searches using the terms “brain tumor” or “glioma” or “meningioma”, “polymorphism*” or “variant*” or “mutation”, “MTHFR” or “MTRR”.Two authors (GXL. and CCH.) independently screened titles and abstracts to identify relevant studies. Published case-control studies will be included in current meta-analysis if they reach the following criteria: (a) evaluating the association between folate metabolism genetic variants (A1298C and A66G) and brain tumor susceptibility in adults; (b) case-control studies on human, and published in English; (c) sufficient data for assessing the ORs and 95%CI, and P value. Exclusion criteria were: (a) case reports, letters, and review articles; (b) containing only case groups; (c) duplication of published articles.Two authors (GXL. and LXC) independently extracted data from all eligible studies. Data such as: (a) the first author name, publication date, country, ethnicity and source of control; (b) cancer types, frequency of cases and controls, involved genes and HWE status in controls. Any disagreements between the two authors were resolved through discussions and agreementsThe strength of association between folate metabolism genetic variants (A1298C and A66G) and meningioma and glioma susceptibility in adults was calculated by odds ratios (ORs) with 95% confidence interval (CI). We used dominant, recessive, homozygote, heterozygote, allelic as the models. Stratified analyses were conducted by cancer type, ethnicity and sources of control. The pooled ORs were calculated for these five models. Heterogeneity was evaluated by using Q statistics (significant at p < 0.05). A Fixed-effect model was utilized if p-value of heterogeneity tests were more than 0.05 (P>0.05) ; otherwi