MTHFR) polymorphisms are implicated in glaucoma remains controversial. Therefore, we performed this study to better assess the relationship between MTHFR polymorphisms and the likelihood of glaucoma.Whether methylenetetrahydrofolate reductase  and 95% confidence intervals (CIs) were calculated.p = 0.04, OR = 0.90, 95%CI 0.81–1.00) and allele  comparisons. Further, subgroup analyses by ethnicity revealed that both rs1801131 and rs1801133 polymorphisms were significantly associated with the likelihood of glaucoma in West Asians. However, no positive results were detected for two investigated polymorphisms in East Asians and Caucasians.A total of 18 studies with 7,168 participants were analyzed. In overall analyses, a significant association with the likelihood of glaucoma was detected for the rs1801133 polymorphism in dominant (Our findings indicated that rs1801131 and rs1801133 polymorphisms may serve as genetic biomarkers of glaucoma in West Asians. PreviouMTHFR rs1801131 (A1298C) and rs1801133 (C677T) polymorphisms with glaucoma, but the results of these studies were conflicting and the sample size of individual studies was relatively small  AND  AND (primary open‐angle glaucoma OR POAG OR primary closed angle glaucoma OR PCAG OR pseudoexfoliation glaucoma OR PXFG OR glaucoma). The reference lists of all retrieved publications were also manually screened to identify other potentially relevant articles.The current meta‐analysis was adhered to the Preferred Reporting Items for Systematic Reviews and Meta‐analyses (PRISMA) guideline  provide adequate data to calculate odds ratios (ORs) and 95% confidence intervals (CIs); (c) full text in English or Chinese available. Studies were excluded if one of the following criteria was fulfilled: (a) not relevant to MTHFR polymorphisms and glaucoma; (b) family‐based association studies; (c) case reports or case series; (d) abstracts, reviews, comments, letters, and conference presentations. For duplicate reports, we only included the study with the largest sample size for analyses.To test the research hypothesis of this meta‐analysis, included studies should meet all the following criteria: (a) case–control study on correlation between 2.3MTHFR polymorphisms in cases and controls. Additionally, the probability value  of Hardy–Weinberg equilibrium (HWE) was also calculated. The Newcastle‐Ottawa scale (NOS) was used to assess the quality of eligible studies  name of first author; (b) year of publication; (c) country and ethnicity of participants; (d) the number of cases and controls; and (e) the genotypic distribution of s Stang, . The NOS2.4MTHFR polymorphisms with the likelihood of glaucoma in all possible genetic models, and a p value of 0.05 or less was considered to be statistically significant. Between‐study heterogeneities were evaluated with I2 statistic. If I2 was greater than 50%, random‐effect models (REMs) would be used to pool the data in overall and subgroup analyses. Otherwise, fixed‐effect models (FEMs) would be employed for synthetic analyses. Sensitivity analyses were carried out to test the stability of our findings. Funnel plots were applied to evaluate possible publication bias.All statistical analyses were conducted with Review Manager Version 5.3.3 . ORs and 95% CIs were calculated to assess potential associations of 33.1The literature search generated 144 results. After exclusion of irrelevant and duplicate articles by reading titles and abstracts, 30 articles were retrieved for further evaluation. Another 12 articles were subsequently excluded after reading the full text. Finally, a total of 18 eligible studies containing 3,453 cases and 3,715 controls were included for analyses  and allele  comparisons. Further subgroup analyses by ethnicity revealed that both rs1801131 and rs1801133 polymorphisms were significantly associated with the likelihood of glaucoma in West Asians. However, no positive results were detected for two investigated polymorphisms in East Asians and Caucasians. When we stratified data based on type of disease, we found that the rs1801133 polymorphism was also significantly associated with the likelihood of primary open‐angle glaucoma (POAG). No any other positive findings were observed in overall and subgroup analyses  and A to C substitution of rs1801131  could lead to reduced enzymatic activity and result in hyperhomocysteinemia, which may partially explain our positive findings (Li et al., MTHFR polymorphisms and glaucoma may be affected by gene–gene and gene–environmental interactions. However, the majority of studies did not consider these potential interactions, which impeded us to perform relevant analyses accordingly (Xie, Shi, & Liu, MTHFR polymorphisms and glaucoma could not be established (Zhao, Yin, Wang, & Si, As with all meta‐analysis, this study certainly has some limitations. First, our findings were based on unadjusted analyses due to lack of raw data, and failure to conduct further adjusted analyses for age, gender, and co‐morbidity conditions may impact the reliability of our findings (Shi, Xie, Jia, & Li, Overall, our meta‐analysis suggested that rs1801131 and rs1801133 polymorphisms may serve as genetic biomarkers of glaucoma in West Asians. However, further well‐designed studies are still warranted to confirm our findings.This article does not contain any studies with human participants or animals performed by any of the authors.None declared.Ling Zhang and Bin Chen conceived of the study, participated in its design, conducted the systematic literature review, performed data analyses, and drafted the manuscript. All authors have read and approved the final manuscript. Click here for additional data file.


EGLN2 and cancer risk has been investigated by several case–control studies in recent years, but investigation results were inconsistent. Thus, a systematic assessment of the association was performed based on a literature review and pooled analysis.The association between a 4‐bp indel polymorphism (rs10680577) within the distal promoter of Two investigators independently retrieved relevant studies from PubMed, Chinese National Knowledge Infrastructure (CNKI), Embase, and Google Scholar. The fixed or random effects model was selected to calculate odds ratios (ORs) with 95% confidence intervals (CIs) based on heterogeneity level. All analyses including heterogeneity assessment, subgroup analysis, sensitivity analysis, and publication bias assessment were performed using RevMan 5.3 software and Stata 12.0 software.EGLN2 rs10680577 polymorphism was significantly associated with cancer risk under all genetic models. However, subgroup analysis based on cancer type showed that the polymorphism was significantly associated with the risk of digestive system cancer under all genetic models, and with the risk of lung cancer under dominant model, heterozygote comparison model, and allele comparison model. Subgroup analysis based on population sources showed a significant association in Chinese population under all genetic models.A total of six relevant studies with 3,406 cases and 5,147 controls were included in the final analysis. The overall pooled analysis showed that EGLN2 rs10680577 polymorphism is associated with cancer risk, and may act as a promising predictive biomarker for cancer risk, especially in Chinese population. However, further well‐designed studies are warranted to confirm these results.The present result suggests that  The last retrieval was updated on 15 January 2019 with the following keywords: “cancer”, “tumor”, “2.2EGLN2 rs10680577 polymorphism and cancer risk; (b) case–control studies; and (c) available genotype frequencies. Meanwhile, the following exclusion criteria were also applied: (a) review, abstracts, case reports, and editorials; (b) studies that did not report genotype frequencies; and (c) studies that reported duplicated results.All articles were reviewed by two investigators independently. Studies were considered eligible if they met the following criteria: (a) investigating the association of 2.3The Newcastle–Ottawa scale was utilized to assess the quality of studies Stang, . A total2.4Two investigators independently extracted data from included studies according to a standardized form. For each study, the following information was extracted: name of first author, publication year, country, cancer type, genotyping method, sample size, and genotype and allele frequencies. Any disagreements will be resolved by discussing with a third investigator.2.5Pz < .05 was considered as the significance threshold for each genetic model. The Chi‐squared test was conducted to evaluate whether these studies deviated from Hardy–Weinberg equilibrium (HWE), and the threshold for disequilibrium was PHWE < .05. Cochran's Q test was performed to assess heterogeneity across individual studies, and HP ≤ .10 suggested heterogeneity. The fixed effects model was selected to estimate the pooled OR if HP > .10; otherwise, the random effects model was adopted. Funnel plots and Egger's test were used to assess the publication bias. EP < .05 indicated significant publication bias.Statistical analysis was performed using Review Manager 5.3 software and Stata 12.0 software. The pooled odds ratios (ORs) with corresponding 95% confidence intervals (95% CIs) were calculated to assess the strength of the association. Both the pooled ORs and the lower limit of 95% CIs > 1 indicated an increased risk. Both the pooled ORs and the upper limit of 95% CIs < 1 indicated a decreased risk. The following five genetic models were used in this meta‐analysis: dominant model [(Ins/Del + Del/Del) vs. Ins/Ins], recessive model [Del/Del vs. (Ins/Del + Ins/Ins)], homozygote comparison model [Del/Del vs. Ins/Ins], heterozygote comparison model [Ins/Del vs. Ins/Ins], and allele comparison model [Del vs. Ins]. A value of 33.1A flow diagram for Literature retrieval strategy is shown in Figure 3.2EGLN2 rs10680577 polymorphism was significantly associated with cancer risk under all genetic models   vs. Ins/Ins:OR = 1.50, 95% CI = 1.35–1.66, ZP < .001; Del/Del vs. (Ins/Del + Ins/Ins): OR = 2.00, 95% CI = 1.53–2.62, PZ < .001; Del/Del vs. Ins/Ins: OR = 2.28, 95% CI = 1.74–2.99, PZ < .001; Ins/Del vs. Ins/Ins: OR = 1.43, 95% CI = 1.28–1.59, PZ < .001; Del vs. Ins: OR = 1.44, 95% CI = 1.32–1.57, PZ < .001], but also with the risk of lung cancer under dominant model , heterozygote comparison model  and allele comparison model . Subgroup analysis based on population sources showed a significant association in Chinese population under all genetic models .As shown in Table 3.3EGLN2 rs10680577 polymorphism and cancer risk under recessive genetic model .Funnel plots and Egger's test were used to assess the publication bias. As shown in Figure 4EGLN2 and cancer risk based on two independent case–control studies, and found that the deletion allele of rs10680577 polymorphism was significantly associated with increased risk of hepatocellular carcinoma. Furthermore, genotype–phenotype correlation studies showed that the deletion allele was significantly correlated with higher expression of EGLN2  a different genetic background between Chinese and Iranian population; (b) a small sample size of the study on Iranian population (only 134 cases and 154 controls); (c) genotype distribution of control samples in Iranian population deviated from HWE.In the year 2012, Zhu et al. firstly investigated the association between a 4‐bp indel polymorphism (rs10680577) within the distal promoter of Compared with previous meta‐analysis, the current meta‐analysis contained more samples and provided more valuable information such as results of subgroup analysis. However, some limitations still existed and needed to be clarified. Firstly, the number of included studies was small and only six case–control studies were analyzed. Secondly, due to insufficient information, potential interactions including gene–gene, gene–environment or gene–some potential covariates were not considered. Thirdly, Literature retrieval strategy was limited by language, and only articles published in English or Chinese were included.EGLN2 rs10680577 polymorphism was associated with cancer risk, and may act as a valuable biomarker for predicting cancer risk, especially in Chinese population. However, further well‐designed studies are warranted to confirm these results.In conclusion, our meta‐analysis determined that the The authors declare no conflict of interest. Click here for additional data file. Click here for additional data file.


Objective: It has been reported that the periodontal disease is linked to a number of malignant tumors such as lung cancer and pancreatic cancer. In this study, we aimed to investigate the association of periodontal disease with risk of bladder cancer by a meta-analysis.Methods: PubMed, Scopus, ScienceDirect, and Chinese National Knowledge Infrastructure (CNKI) were searched for eligible publications up to December 15, 2017. Cohort and nested case-control studies on the association between periodontal disease and risk of bladder cancer were included. After study selection and data extraction, pooled hazard ratios (HRs) and their 95% confidence intervals (95%CIs) were calculated using a fixed-effect inverse-variance model. All analyses were performed using the RevMan 5.3 software.Results: Finally, five cohort studies were identified and included in this meta-analysis, involving 1,104 bladder cancer cases of 298,476 participants. Summary estimates based on adjusted data showed that periodontal disease was not significantly associated with the risk of bladder cancer . A similar result was also observed after cumulative, subgroup and sensitivity analyses.Conclusions: Current evidence from cohort studies suggests that patients with periodontal disease may not be at an increased risk of developing bladder cancer. According to the 2015 Global Burden of Disease (GBD) study, the cumulative burden of oral conditions including untreated dental caries, severe chronic periodontitis and total tooth loss had dramatically increased between 1990 and 2015, as accounting for a 64% increase in disability-adjusted life years (DALYs); and the incident cases in 2015 were 616 million , the studies meeting each of the following criteria were included in the meta-analysis: (1) the study (S) adopted a prospective cohort, retrospective cohort, or nest case-control design and provided full text; (2) the disease (P) was BC, the exposure (I) of interest was PD and the control (C) was free of PD; (3) the outcome (O) of interest was incident BC ; (4) adjusted risk ratios (RRs), incidence density ratios, or hazard ratios (HRs) with their associated 95% confidence intervals (95% CIs), and the adjustment for confounding factors were reported. If more than one reports came from the same cohort, we analyzed them and then chose the best one.We searched PubMed, Scopus, ScienceDirect and Chinese National Knowledge Infrastructure (CNKI) to identify all relevant studies published in English or Chinese up to December 15, 2017 using the following keywords: periodontal disease, periodontitis, and cancer. Reference lists of included studies as well as the newest reviews and editorials were also screened for additional studies.All the studies were selected according to the aforementioned inclusion criteria independently by two authors. Then the following information was extracted from each eligible study: first author's surname, publication year, country of origin, name of cohort, sample size, cases of BC, gender, age, smoking status, outcome measures, data on adjusted HRs/RRs and their 95% CIs, and covariates for adjustment in the multivariable model. Any discrepancies were resolved by discussion or by consulting a third author.Q-test  and the I2 statistic (I2 ≥ 50% suggesting significant heterogeneity) . Incidence density ratios and RRs were considered as HRs in meta-analyses , all other subgroups with no or non-significant homogeneity. The retrospective cohort, male and female, and Asia subgroups with OR < 1.00; all others were >1.00. Sensitivity analyses by omitting the study with longest follow-up duration, study with shortest follow-up duration, co-twin study, or the study without adjustment of smoking status, all indicated that the overall result was robust and with good homogeneity.Table Growing evidence suggests that PD has common risk factors with a number of other non-communicable diseases and conditions, and the exploration of the common behavioral and environmental risk factors may contribute to the effective prevention of PD. In this meta-analysis, the relationship between PD and BC was evaluated by incorporating five cohort studies with 1,104 BC cases and 298,476 samples, and a positive but non-significant result was obtained. The cumulative, subgroup and sensitivity analyses all supported that this non-significant result was not influenced by type of study design, follow-up time, ethnicity, smoking and alcohol status, or gender.Our results are similar to that of another meta-analysis on colorectal cancer, which indicated null associations between PD and specific types of cancers. However, evidence from other studies that reported positive association between PD and cancers with positive results indicating that PD might play an important role in the development of certain cancers (Sadighi Shamami and Amini, It's biologically plausible that smoking and PD act jointly to increase the risk of cancer (Zeng et al., Several limitations existed in observational studies on PD and cancer risk (Bueno et al., In conclusion, this meta-analysis revealed that PD might not be associated with risk of BC, which was not affected by study design, duration of follow-up, ethnicity, smoking and alcohol status, or gender. However, we could not downplay this issue, because PD is implicated in the occurrence mechanisms of many systematic diseases, and BC patients with PD also need early aggressive treatment to prevent other diseases. Besides, further relevant studies should adopt standardized measurements and definitions for categories of PD, thereby making the results more comparable and complete.W-ZX, X-HW, and X-TZ designed this study. W-ZX and Y-HJ performed search and collected data. W-DL re-checked data. Y-HJ and X-TZ performed analysis. X-TZ wrote the manuscript. W-DL and X-HW reviewed the manuscript.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.


Vitamin D deficiency and frailty are common with aging. Previous studies examining vitamin D status and frailty have produced mixed results, and in particular, the shape of the association has not been well established. We examined the association between 25-hydroxyvitamin D (25OHD) serum levels and frailty by performing a systematic review and dose-response meta-analysis.We searched the PubMed, EMBASE and Cochrane Library databases of Elsevier through February 2017. Cross-sectional and cohort studies that reported adjusted risk ratios with 95% confidence intervals (CI) for frailty with ≥3 categories of 25OHD serum levels were selected. Data extraction was performed independently by two authors. The reported risk estimates for 25OHD categories were recalculated, employing a comprehensive trend estimation from summarized dose-response data.I2 = 86.8%) in the 6 cross-sectional studies and 0.89  in the 4 prospective cohort studies. Based on the Akaike information criteria (AIC), a linear model was selected .The pooled risk estimate of frailty syndrome per 25 nmol/L increment in serum 25OHD concentration was 0.88  contains supplementary material, which is available to authorized users. With increasing age, blood vitamin D concentrations decrease due to decreased kidney function, diminished sun exposure, intrinsic skin response to ultraviolet radiation and poor diet . VitaminFrailty is a clinical state in which an individual’s vulnerability to developing increased dependency and/or mortality when exposed to a stressor is increased –7. NumerAn increasing number of studies investigating the association between 25-hydroxyvitamin D (25OHD) and frailty have yielded conflicting information. Although hypovitaminosis D can potentially increase the risk of frailty, not all observational studies have confirmed this relationship , 12. EviFurthermore, the exact relationships, including whether a dose-response pattern exists, are currently unclear. Defining which levels of 25OHD are strongly associated with frailty syndrome is important for shaping elderly health recommendations about vitamin D supplementation considering the optimal serum 25OHD concentration. Furthermore, a recent prospective cohort study  in commuWe searched the PubMed, Cochrane Library, and EMBASE databases via Elsevier through February 2017. A medical librarian together with the reviewers developed database-specific search strategies according to the particular subject headings and searching structure of the databases  an observational design including cross-sectional studies and cohort studies in humans, 2) the inclusion of frailty as a specified outcome, 3) a baseline assessment of serum 25OHD levels, 4) the inclusion of data on relative risk (RR) and its corresponding 95% confidence interval (CI) or data to calculate these values for frailty syndrome for each category of serum 25OHD level, and 5) the inclusion of the most recent and complete study  if cohorts were duplicated in more than one study.Review articles, editorials, commentaries, and letters with no new data analysis, meta-analyses, and abstracts were excluded. The exclusion criteria for this study were as follows: 1) an experimental design was used, 2) the outcome was not frailty, and 3) only two serum 25OHD levels were specified. Two investigators (Sang Yhun Ju and Do Hoon Kim), coauthors of the present study, independently extracted the data from the original reports. The following information was extracted: the first author’s family name, year of publication, country of origin, the mean or median age of the participants, gender, sample size, the number of participants for each serum 25OHD level, the number of cases for each serum 25OHD concentration category, adjusted covariates, definitions of frailty used, the method of 25OHD assessment, follow-up duration, and categories of serum 25OHD and their corresponding RRs with their 95% CIs for frailty. The adjusted risk estimates that reflected the most comprehensive control were extracted to avoid potential confounding variables. Disagreements between the two reviewers were resolved by consensus. We planned, conducted, and reported this systematic review according to the widely accepted quality standards .The methodology of the statistical analysis has been described in detail elsewhere . In brieatistics . We rega’s tests  and visu’s tests  to identThe process of identifying and selecting the studies is summarized in Fig. Table The reported risk estimates for the association between 25OHD level intervals and frailty are illustrated in Fig. I2 = 86.8%) in the six cross-sectional studies and 0.89  in the four prospective cohort studies , but not for the prospective cohort studies . The trim-and-fill sensitivity method imputed estimates from three hypothesized negative unpublished estimates. The corrected effect estimate for a 25-nmol/L increase in the 25OHD level was reduced to 0.98 , demonstrating no relationship between the risk of frailty and serum 25OHD levels after accounting for a potential publication bias.For comparisons with results from other studies and reviews, the RRs for the 50 and 75 nmol/L increases in 25OHD levels were also estimated: 0.65 (95% CI = 0.50–0.85) and 0.52 (95% CI = 0.35–0.78), respectively, for cross-sectional studies and 0.76 (95% CI = 0.66–0.87) and 0.66 (0.53–0.82), respectively, for prospective cohort studies. The funnel plot of the linear dose-response slopes was somewhat asymmetric for the cross-sectional studies, with smaller studies tending to have larger risk estimates, suggesting publication bias. A publication bias was detected with Egger’s test for the cross-sectional studies , the subgroup (nine frailty indicators and the frailty index) of frailty definition , the subgroup of other 25OHD measurement methods  and the Newcastle-Ottawa scale ≥7 subgroup . In addition, within the prospective cohort studies, we found that the association was significant for phenotype definition of frailty , duration of follow-up ≥5.6 year  and the subgroup (yes) of key-set covariates .Subgroup analyses were performed based on geographic location , Newcastle-Ottawa Scale (≥7 vs < 7), and follow-up years (≥median vs. <median). The subgroups are presented in Table To our knowledge, this is the first dose-response quantitative systematic review of observational studies investing the effect of serum 25OHD levels on the risk of frailty using data from both cross-sectional and prospective cohort studies. This meta-analysis of data from more than 20,000 study participants demonstrates a statistically significant inverse association between serum 25OHD levels and the risk of frailty, and this finding was consistent across subgroups. A 25-nmol/L increase in 25OHD levels was associated with an 11% decrease in the incidence of frailty in prospective cohort studies and a 12% decrease in the risk of frailty in the cross-sectional studies. A statistically linear relationship between serum 25OHD levels and the risk of frailty was also found, even after adjustments for other known risk factors.We estimated a protective effect of a 25-nmol/L increase in 25OHD levels against frailty because this number is the difference between the upper limit of vitamin D deficiency, defined by most experts as a 25OHD level less than 50 nmol/L, and the lower limit of sufficient vitamin D levels above 75 nmol/L. Additionally, a daily intake of at least 1000 IU  of vitamin D3 appears to be required to elevate vitamin D concentrations by 25 nmol/L, which would ensure that no less than 50% of the population has the recommended 25OHD level of at least 75 nmol/L . SeveralA previous meta-analysis  of sevenOur results suggest that high 25OHD levels are associated with a lower risk of frailty in elderly people. Conversely, frailty itself may contribute to lower 25OHD by reducing the levels of outdoor activity and sunlight exposure. Elderly individuals with frailty are at a high risk of developing vitamin D deficiency due to decreased dietary intake, less sun exposure, and a decreased capacity to produce sufficient amounts of calcitriol due to an age-related decline in hydroxylation by the kidney . The cauConsiderable overlap exists between sarcopenia and frailty, especially in terms of the physical aspects of the frailty phenotype: low grip strength, gait speed and muscle mass . While tThe last pathway through which low vitamin D may affect frailty is related to its hypothesized anti-inflammatory properties . SeveralThe findings of our study should be interpreted within its limitations. The included studies have no data regarding vitamin D supplementations evaluated through food frequency questionnaires or self-administered questionnaires, which could affect serum 25OHD concentrations; therefore, the present study may under- or overestimate our results. The definitions for frailty used in the included studies were different , thus affecting our pooled analysis, although subgroup analysis was performed according to definitions of frailty. Because several eligible studies did not provide sufficient information for a dose-response analysis of 25OHD levels, the number of participants, cases, and logarithms of RRs and corresponding standard errors, we excluded the potential related studies , 45–49, Our findings suggest that 25OHD serum levels are independently associated with the risk of frailty, which is consistent with the results of a nonlinear analysis and a linear regression analysis. Further interventional research should investigate whether vitamin D supplementation can be useful for preventing frailty in the elderly population.Additional file 1:Search strategy. (DOCX 24 kb)Additional file 2:MOOSE Checklist for Meta-analyses of Observational Studies. (DOCX 55 kb)Additional file 3:Quality of the observational studies in the meta-analysis based on the Newcastle-Ottawa Scale. (DOCX 14 kb)Additional file 4:Figure S1. Begg’s Funnel plot with 95% confidence intervals in the meta-analysis of the cross-sectional studies. Figure S2. Begg’s Funnel plot with 95% confidence intervals in the meta-analysis of the prospective cohort studies. (DOCX 59 kb)Publication bias for association between serum 25-hydroxyvitamin D concentration per 25-nmol/L increment and frailty syndrome.


The studies included in this review and meta-analysis revealed that dietary fats and fatty acids had no effects on the risk of CRC.Dietary fat intake is associated with the risk of colorectal cancer (CRC); however, the results of epidemiological studies on this are controversial. Therefore, this study aimed to summarize the available scientific evidence regarding the association between dietary fat and the risk of CRC. We conducted a systematic search of PubMed, Web of Science, and the Cochrane library for articles related to dietary fat and the risk of CRC. The summary relative risks with 95% confidence intervals (CI) were calculated via a random effect model. Begg’s test was used to detect publication bias. A total of 18 articles were identified. The pooled relative risk with 95% CI for the risk of CRC were 1.00 (95% CI: 0.90–1.12), 0.97 (95% CI: 0.86–1.10), 1.08 (95% CI: 0.92–1.26), and 0.99 (95% CI: 0.93–1.04) for total fat, saturated fatty acid, monounsaturated fatty acid, and polyunsaturated fatty acid, respectively. No significant associations were found in subgroup analyses. Begg’s test for all exposures revealed no publication bias (total fat,  Colorectal cancer (CRC) is one of the most common cancers in both men and women worldwide ,2,3. TheAmong factors associated with the development of CRC, modifiable risk factors, such as physical activity, smoking, alcohol drinking, and diet, have drawn attention ,11. In pAccordingly, the need for meta-analyses that qualitatively assess individual studies to integrate recent literature on the association between total fat and individual fatty acid intake and the risk of CRC is evident. Therefore, in this study, we reviewed the available literature and summarized the results by conducting a meta-analysis of prospective cohort studies.A systematic literature review and meta-analysis of studies on the association between dietary fat intake and the risk of CRC were conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guideline . PubMed,Duplicates were removed using the ENDNOTE program. Titles and abstracts were reviewed to check whether the selected articles satisfied the inclusion criteria. If the eligibility for inclusion could not be determined based on the title and abstract only, the full text was reviewed for the final inclusion. Two investigators performed the selection process independently, and differences in opinion between investigators regarding study selection were resolved by discussion.The inclusion criteria were as follows: (1) Original research, (2) prospective cohort study, (3) dietary fat as the major exposure factor ), (4) CRC as the major outcome , and (5) relative risk (RR) or hazard ratio (HR) and the 95% confidence interval (CI) were reported. The exclusion criteria were as follows: (1) Papers other than an original research  and (2) not conducted in humans . For studies overlapping in studied populations, the most recently published article was included. The following information was extracted by reviewing the full texts of the selected articles: Name of authors, year of publication, country, study population, sex, age at baseline, total number of participants, follow-up period, method of diet assessment, diagnostic criteria of CRC, intake levels of fat or fatty acids, RR/HR and 95% CI, and confounders. Quality evaluation was performed according to the Newcastle-Ottawa quality assessment scale-cohort studies , which c2 tests were used to evaluate the heterogeneity of studies included in this analysis. Studies were deemed heterogeneous if I2 was >50% or the p value in the Cochrane’s Q test was <0.1 [p <0.05.To combine the effect size between dietary fat intake and the risk of CRC from each study, the RR or HR and 95% CI of the group with the highest fat intake relative to those of the group with the lowest fat intake were used. The weighted values needed to combine the effect sizes were calculated by identifying the standard deviation of a log-transformed RR, and the summary RR was determined using a random effect model. Cochrane’s Q and Higgin’s Iwas <0.1 . Subgroun = 8156), articles lacking information (n = 44), articles which were not original studies (n = 4065), articles not involving human participants (n = 8787), and articles unrelated to the topic of this study (n = 7288) were excluded. Among the 66 remaining studies, those with overlapping datasets (n = 6), those without RR or HR and 95% CI (n = 3), and those that were not prospective cohort studies (n = 39) were excluded. Overall, 18 articles were included in this meta-analysis , and no significant association was also observed . Similarly, the pooled results of the seven articles reporting the MUFA-CRC association were not significant . The results of the 14 articles on PUFA, including omega-3 and omega-6 fatty acids, were pooled, and no significant association between PUFA intake and risk of CRC was found .In the subgroup analysis of the association between total fat intake and CRC according to the continent in which the study was conducted, heterogeneity was reduced in the United States and Europe, but the association was not significant . SimilarTo verify the reliability of these results, we conducted the leave-one-out analysis to assess the influence of each study on the summary RR. When the study by Chyou et al., (including men) was excluded , the assp = 0.3). Moreover, SFA showed an asymmetrical trend at the lower right side of the funnel plot; however, Begg’s test also showed no publication bias (p = 0.1). Similarly, no publication bias was observed on the funnel plots, and Begg’s test results for MUFA and PUFA also showed no publication bias . Publication bias was visually analyzed using the funnel plot, and the test statistics were determined using Begg’s and Mazumdar’s rank correlation test. The funnel plot for total fat intake showed an asymmetrical trend, but Begg’s test showed no publication bias  and risk of CRC showed a positive association ,43. The In our study, we found that PUFA intake was not associated with CRC. Some studies have reported a positive association between levels of PUFA intake and CRC ,50, wherAlthough multivariable-adjusted RRs were reported in individual studies, the range of adjustment varied across studies, with the possibility of residual confounding. However, the possibility of reverse causality or information bias was minimized by integrating results of the prospective cohort study. Moreover, unlike previous studies, the current study assessed the quality of articles included in the meta-analysis to prevent the distortion of research results by poorly designed studies or from small study effects. Finally, because no meta-analyses integrated the research results on this topic since 2011, this study included studies published after 2011 to enhance the evidence level on the unclear association between dietary fat intake and CRC. This study systematically reviewed research articles published since the 1990s and integrated the results to assess the influence of dietary fat or fatty acids on the risk of CRC. No significant association was observed between total fat, SFA, MUFA, and PUFA intakes and the risk of CRC. Moreover, subgroup analyses by sex, continent, and follow-up period revealed no significant differences. Because numerous studies have been published in the United States and Europe, large-scale prospective studies and clinical trials involving a greater diversity of countries and races should be conducted.


However, rs352140 was not related to cervical cancer regardless of whether the subgroup analysis was conducted (p > 0.05). In conclusion, there is a significant correlation between rs187084 and cervical cancer risk with the minor C allele increasing the risk of occurrence of cervical cancer. However, rs352140 is not associated with the occurrence of cervical cancer.This meta-analysis systematically reviews the association between Toll-like receptor 9 polymorphisms and the risk of cervical cancer. Case-control studies focused on the association were collected from the PubMed, Web of Science, Cochrane Library, Embase, MEDLINE, CNKI, VIP, and Wanfang databases from inception to July 2017. We screened the studies and assessed the methodological quality of the included studies and extracted data. A meta-analysis was performed using RevMan 5.3 and Stata 12.0 software. Pooled odds ratios and 95% confidence intervals were employed to evaluate the strength of the associations between Toll-like receptor 9 polymorphisms and cervical cancer risk. A total of 9 studies comprising 3331 cervical cancer patients and 4109 healthy controls met the inclusion criteria. Of these, 8 studies contained information about G2848A (rs352140) and 4 studies contained information about −1486T/C (rs187084). Our results revealed that the associations between rs187084 and cervical cancer risk in the dominant model ( Cervical cancer is the fourth most common cancer in women in terms of both incidence and mortality worldwide . AccordiTLR9 expression increases according to the histopathological grade of the cervical pathological process and the HPV E6 and E7 oncoproteins deregulate the expression and function of TLR9 /cervix cancer/cervical cancer, and the combinations of these. Additionally, we searched the reference lists of all identified articles manually for more studies.We reviewed the PubMed, Web of Science, the Cochrane Library, Embase, MEDLINE, CNKI, VIP, and Wanfang databases systematically and comprehensively. The search terms are the following: Toll-Like Receptors [Mesh] or TLRTLR9 gene polymorphisms (rs352140 and rs187084) and the risk of cervical cancer, (2) human study subjects, (3) case-control studies, (4) available and sufficient genotype distribution of data to calculate odds ratios (ORs) and corresponding 95% confidence intervals (CIs), and (5) diagnoses based on cervical biopsy pathology. Additionally, if there were duplicate studies, we made sure that the most recent or the most complete one was included. If it did not satisfy the criteria above, the article was excluded.Studies included needed to meet the following criteria: (1) a focus on the association between the Two investigators extracted relevant data from all the eligible studies independently. A third reviewer was invited to participate in the work when some disagreement occurred; consensus was ultimately reached by discussion. These characteristics were collected from each study: the first author, publication year, race, total numbers of cases and controls, study design, source of controls, genotyping method, and evidence of HWE in controls.The quality of included studies was assessed using the Newcastle-Ottawa scale, including three categories: selection, comparability, and outcome. Additional eight items were used to assess the methodology of each qualified study. The highest score was 9. Studies with a score of more than 7 were considered as high quality. A study awarded a score of 0–3, 4–6, or 7–9 was considered as a low-, moderate-, or high-quality study, respectively.Z-test revealed statistical significance when p < 0.05. I2 and Q statistics were employed to detect heterogeneity among different studies. There was no heterogeneity if I2 < 50% and p > 0.1 and a fixed effects model was used; otherwise, we thought that heterogeneity existed in the incorporated populations and a random effects model was used instead. Subsequently, we conducted a subgroup analysis according to race. Hardy-Weinberg equilibrium (HWE) was evaluated by χ2 test in control groups with p < 0.05 indicating a deviation from HWE. Sensitivity analysis was utilized to estimate the robustness and stability of the meta-analysis results by deleting all the studies one by one. Next, Begg's funnel plot and Egger's test were used to evaluate publication bias. For each SNP, five genetic models were evaluated to assess the correlation with cervical cancer susceptibility: the allele model, dominant model, recessive model, heterozygote model, and homozygous model. The statistical analyses were performed using RevMan 5.3 and Stata 12.0 software. All p values were two sided, and p < 0.05 was considered to be statistically significant.The odds ratios (ORs) and 95% confidence interval (CIs) were applied to assess the strength of the correlation between SNPs and cervical cancer susceptibility. A TLR9 polymorphisms, 4 were review papers, and 2 were meta-analyses. We downloaded the remaining 10 articles as full-text reports and reviewed them carefully. One record was excluded for containing duplicate samples. Finally, 9 case-control studies containing 3331 cases and 4109 controls were included in the meta-analysis, among which 8 studies were about rs352140 (G2848A) and 4 articles were about rs187084 (−1486T/C). All studies were based on Caucasian or Chinese Han populations.By searching the electronic databases systematically, we initially retrieved 72 articles . After eThese 9 case-control studies were published between 2011 and 2017. Four studies were performed with Caucasians, while five papers were based on Chinese Han populations. All the studies were performed using PCR-RFLP to identify the polymorphism sites except two studies, which used the TaqMan and Illumina GoldenGate methods. Two of the nine control groups were hospital-based groups, while the rest were population-based groups .We applied the Newcastle-Ottawa scale (NOS) to estimate the quality of the nine included studies . The resp = 0.09) (p = 0.08), recessive model , heterozygote model , or homozygous genetic model  (p > 0.05), but the A allele was correlated to increasing cervical cancer susceptibility in Caucasians based on the allele genetic model   , dominan = 0.23) . Conside = 0.03) . After ep = 0.02), dominant model  (p = 0.002); nevertheless, it was not related to cancer susceptibility on the recessive model or homozygous genetic model (p > 0.05)  (p = 0.02). There was no statistical correlation in the recessive model  or homozygous genetic model   , and het > 0.05) . We then = 0.05)  and hete = 0.67) , Table 4p = 0.02), dominant model  (p = 0.002). However, the SNP was not related to cancer susceptibility in the recessive model or homozygous genetic model (p > 0.05)  (p = 0.02). There was no statistical correlation in the recessive model  or homozygous genetic model   , and het > 0.05) . We perf = 0.05)  and hete = 0.67) , Table 4I2 > 50%, p < 0.1). In consideration of this, we employed a random effects model for the meta-analysis. Additionally, the subgroup analysis was stratified by race to eliminate heterogeneity , dominant model , recessive model , heterozygote model , and homozygous genetic model . This implied that race might be a confounding factor and source of heterogeneity, whereas the pooled ORs were robust substantially.As presented in ogeneity . It was I2 = 43%, p = 0.15), dominant model , or heterozygous genetic model . However, heterogeneity was present in the recessive model  and homozygous genetic model . Similarly, a random effects model and subgroup analysis were conducted to evaluate heterogeneity. We found that the heterogeneity decreased in the Chinese Han population based on the homozygous genetic model   , but no  = 0.24) .p < 0.05). However, the results did not change substantially after removing them. Furthermore, a sensitivity analysis was utilized to evaluate the stability of the meta-analysis by deleting all the studies one by one. However, the overall ORs did not change significantly in any of the genetic models, indicating that the meta-analysis was robust and stable. The sensitivity analysis of the associations between the two SNPs and cervical cancer susceptibility based on the dominant genetic model is shown in Figures As mentioned before, two studies focused on rs352140 and one focused on rs187084 were not consistent with the balance of HWE in the control groups (p > 0.05)  while the pooled results in other genetic models remained steady. The evaluated FPRP was 0.751, which was greater than 0.2, implying a false positive association between rs352140 and cervical cancer risk among the Caucasian population [Subgroup analysis of rs352140 showed a significant decline in heterogeneity in the Caucasian population, implying that the heterogeneity may result from racial differences. Moreover, it was statistically significant that the A allele elevated the cervical cancer risk in the allele genetic model  and cervical cancer susceptibility, while rs187084 (−1486T/C) has a statistical correlation with the risk of cervical cancer. The rs187084 C allele is considered a deleterious allele and increases cervical cancer risk. Further large, well-designed case-control studies are needed to authenticate these results.In conclusion, this meta-analysis demonstrated that there is no association between


TNF-α polymorphisms -308G/A and -238G/A and T2DM remains controversial. The present study sought to verify associations between these polymorphisms and T2DM susceptibility using a meta-analysis approach. A total of 49 case–control studies were selected up to October 2018. Statistical analyses were performed by STATA 15.0 software. The odds ratios (ORs) and 95% confidence intervals were calculated to estimate associations. Meta-analyses revealed significant associations between TNF-α −308G/A and T2DM in the allele model (P=0.000); the dominant model (P=0.000); the recessive model (P=0.001); the overdominant model (P=0.008) and the codominant model (P=0.000). Subgroup analyses also showed associations in the allele model (P=0.006); the dominant model (P=0.004) and the overdominant model (P=0.005) in the Caucasian and in the allele model (P=0.007); the dominant model (P=0.014); the recessive model (P=0.000) and the codominant model (P=0.000) in the Asian. There were no associations between TNF-α −238G/A and T2DM in the overall and subgroup populations. Meta-regression, sensitivity analysis and publication bias analysis confirmed that results and data were statistically robust. Our meta-analysis suggests that TNF-α −308G/A is a risk factor for T2DM in Caucasian and Asian populations. It also indicates that TNF-α −238G/A may not be a risk factor for T2DM. More comprehensive studies will be required to confirm these associations.Tumor necrosis factor-α (TNF-α) is involved in insulin resistance and has long been a candidate gene implicated in type 2 diabetes mellitus (T2DM), however the association between  Individuals with type 2 diabetes mellitus (T2DM) accounted for 90% of this total [Diabetes is a global epidemic, with an estimated worldwide prevalence of 1 in 11 adults (approximately 425 million people in 2017), and is projected to increase to 629 million people by 2045  induces insulin resistance to some extent, through the inhibition of intracellular signaling from the insulin receptor [TNF-α promoter polymorphisms −308G/A and −238G/A have been studied in T2DM etiology [Some receptor ,5. The dreceptor . Severalreceptor . Of multetiology .TNF-α promoter lead to T2DM susceptibility. Two large-scale British association analyses found these polymorphisms were not robustly associated with T2DM [Currently, it is inconclusive whether these polymorphisms (−308G/A and −238G/A) in the ith T2DM ,12 and sith T2DM ,14 and Iith T2DM . Howeverith T2DM  and Iranith T2DM . StudiesThis meta-analysis was conducted according to the recommendations of the Preferred Reporting Items for Systematic Reviews and Meta-Analysis 2009 (PRISMA2009). All published studies up to October 2018 were searched using the PubMed, Embase, EBSCO, OVID, and Web of science database. We used the following terms: ‘TNF-α’, ‘TNF-alpha’, ‘tumor necrosis factor-α’, ‘tumor necrosis factor-alpha’, ‘T2DM’, ‘type 2 diabetes mellitus’, ‘type 2 diabetes’, ‘type II diabetes’, ‘non-insulin dependent diabetes’, ‘NIDDM’, ‘polymorphism’, ‘variation’, ‘−308G/A’, ‘rs1800629’, ‘−238G/A’ and ‘rs361525’. Relevant references in selected articles were also included.TNF-α polymorphisms (−308G/A and −238G/A) with the risk of T2DM, (2) the study was case–control designed, (3) sufficient information on genotype frequencies  in both cases and controls to estimate an odds ratio (OR) with a 95% confidence interval (95% CI), (4) all data were original. Exclusion criteria were as follows: (1) other DM (diabetes) types were excluded, (2) non-human studies, (3) reviews, meta-analysis and non-case–control studies and (4) studies not published in English.All articles were independently reviewed by two investigators. Studies were assessed against the following inclusion criteria: (1) the associated study of Study quality was assessed to guarantee the strength of results and conclusions. Quality assessment was performed according to the Newcastle–Ottawa Quality Assessment Scale (NOS), which is a validated scale for nonrandomized studies in meta-analyses . This NOData were independently extracted by two investigators using a standardized form. For each study, the following information was extracted: (1) name of first author; (2) year of publication; (3) ethnicity of population; (4) sample sizes and genotype distributions; (5) allele frequency of the major variant. Ethnicity was categorized as Caucasian, Asian and African.P<0.05. STATA  software was used to calculate meta-analysis results. Individual study heterogeneity was assessed by Cochran’s Q test and the 2I statistic (P<0.10 and 2I > 50% indicates evidence of heterogeneity) [TNF-α promoter polymorphisms (−308G/A and −238G/A) and T2DM risks. Five genetic models were used in this meta-analysis: (1) the allele model ; (2) the dominant model (GA+AA vs. GG); (3) the recessive model (AA vs. GA+GG); (4) the codominant model  and (5) the overdominant model (GG+AA vs. GA). A P-value <0.05 was accepted as the significant threshold for each genetic model. Three subgroups, including Caucasian, Asian and African, based on ethnicity, were analyzed to reduce influences from genetic backgrounds. A meta-regression was used to search the source of heterogeneity [2 resI explained the proportion of residual variation due to heterogeneity, and 2adj R explained the proportion of between-study variation due to heterogeneity [2 resI close to 100% and 2adj R close to 0% further indicated no effects on heterogeneity. Pooled estimates were performed to sensitivity analysis which involved omitting one study at a time followed by recalculation to test for robustness of the summary effects [P-value <0.05 was accepted as statistically significant.The Hardy–Weinberg equilibrium (HWE) test was calculated using the Chi-squared test. The distribution of allele frequencies in controls was considered to deviate from HWE when geneity) . The fixgeneity) , when thgeneity) ,21. ORs ogeneity , which cogeneity . I2 res ogeneity ,26. An I effects . A P-valTNF-α −308G/A and −238G/A. Through in-depth full-text analysis of the remaining 211 publications, 49 publications were used for the final meta-analysis ; the dominant model ; the recessive model ; the overdominant model ; and the codominant model . TNF-α −238G/A was not associated (P>0.05) with T2DM in all genetic models ; the dominant model ; the overdominant model , and also in Asian populations in the allele model ; the dominant model ; the recessive model ; the codominant model  and no associations between −308G/A and T2DM risk in African populations (P>0.05). For −238G/A, it was not associated (P>0.05) with T2DM in the subgroup population , sample size , HWE  and ethnicity, including Caucasian (P=0.106), Asian (P=0.127), using the 10000 times Monte Carlo permutation test. The −238G/A results revealed no influence from publication year (P=0.573), sample size (P=0.498) and ethnicity, including Caucasian (P=0.864) and Asian (P=0.735), using the 10000 times Monte Carlo permutation test. Sensitivity analysis revealed that some studies [P=0.011).The following covariates were considered for meta-regression: publication year, sample size, ethnicity and HWE in controls. The −308G/A results revealed no influence on the publication year . The Begg’s and Egger’s tests showed no existing publication bias in the overall population for all genetic models . Finally, we performed sensitivity analysis to explore the impact of a single study; our results revealed that the study by Golshani et al. [To prevent possible interference from heterogeneity to our results, we sought to explain the source of heterogeneity and eliminate it. First, subgroup analysis of ethnicity and genetic models reduced between-study heterogeneity. We found that heterogeneity was reduced, but there was still high heterogeneity. Next, our meta-regression analysis attempted to reveal these heterogeneous sources. These results showed that publication year, sample size, ethnicity  and HWE were not the sources of between-study heterogeneity  were not included due to insufficient information from selected publications.TNF-α −308G/A were associated with T2DM susceptibility. Additionally, we found that TNF-α −238G/A is not associated with T2DM in overall and subgroup populations. In the future, the influences of genetic loci, combined with environmental factors, may provide important treatment therapies for T2DM, therefore, well-conceived studies are warranted to confirm the important data presented here.In conclusion, our meta-analysis identified that


MMP-8 variants and cancer risk, but they have been unsuccessful in drawing reliable conclusions.Several studies have focused on the relationship between MMP-8 C-799 T, Lys460Thr, and Lys87Glu polymorphisms and cancer risk. We further employed in silico tools to evaluate the effect of MMP-8 expression on cancer susceptibility and overall survival time.We employed odds ratio (OR) together with 95% confidence interval (CI) to assess the correlation between MMP-8 variants and cancer susceptibility was not significant . Similar results were observed in subgroup analysis by ethnicity, cancer type, and source of control. In silico analysis indicated that MMP-8 expression was elevated in bladder cancer tissue compared to that in the control. However, both the higher and lower MMP-8 expression groups did not show an impact on the overall survival time of the patients.A total of 8140 patients with malignant carcinoma and 10,529 healthy individuals (control) were enrolled. Overall, the analysis showed that the relationship between three MMP-8 C-799 T, Lys460Thr, and Lys87Glu variants are not participant with the susceptibility of cancer. It is well-known that the development of carcinoma is complex and has not been completely clarified. Hereditary material and genetic polymorphisms may probably have an impact on cancer susceptibility and play a crucial role in the tumorigenesis of numerous carcinomas –4. PreviMMP-8 gene comprises twelve exons and is located on chromosome 11q22.3 [MMP8 can lead to gene dysfunction, microenvironment disorder and potential carcinogenesis. It has been reported that several single nucleotide polymorphisms (SNPs) of MMP8 can influence the gene expression by altering its promoter activity. In addition, the T allele of MMP8 C-799 T variant was reported to be related to breast carcinoma susceptibility and lymph node metastasis among Asian and Caucasian population [MMP8 C-799 T variation [MMP8 C-799 T was not associated with urinary bladder cancer susceptibility in Caucasian descendants [Previous studies have showed evidence that genetic mutations and variants can predispose for malignant tumors , 19. Hum 11q22.3 . Polymorpulation . Furthercendants .MMP8 genetic variants including C-799 T (rs11225395 C/T), Lys460Thr (rs35866072 A/C), and Lys87Glu (rs1940475 G/A) SNPs, with cancer risk. Research on these MMP8 variations have been carried out in numerous countries, such as USA, Mexico, Poland, India, Korea, and China [MMP8 variations and various cancers among different case control studies. Therefore, we conducted a comprehensive analysis based on accumulated data of all eligible studies to investigate the impact of MMP-8 C-799 T, Lys460Thr, and Lys87Glu polymorphisms on overall cancer susceptibility.A number of studies have evaluated the association of nd China –35. NeveComprehensive literature search on PubMed, Web of Science, EMBASE (Excerpta Medica Database), and SinoMed (China Wanfang Databases) was carried out to identify all eligible case-control studies, prior to June 2019. Valid keyword search strings are as follows:  AND (polymorphism OR variant OR mutation OR SNP) AND . Furthermore, we independently retrieved the references in the identified articles to screen other available studies, with no language restriction. For studies with overlapping data, we selected the most recently published ones.MMP-8 C-799 T, Lys460Thr, and Lys87Glu polymorphisms and cancer risk; (b) case-control study; and (c) contain available data on the frequency of genotypes. We independently excluded unpublished case reports, letters, reviews, meta-analyses, or missing genotype data for C-799 T, Lys460Thr, and Lys87Glu variants. We also excluded the studies that focused only on the case population.Eligible studies should meet the following inclusion criteria: (a) evaluate the relationship between MMP-8 C-799 T, Lys460Thr, and Lys87Glu variants in cases and controls, P-value of Hardy-Weinberg equilibrium (HWE) in controls. The quality score of the eligible studies was evaluated by Newcastle-Ottawa Scale (NOS). The research was regarded as high-quality if it acquired six or more stars.Two authors (LFZ and YYM) independently completed the data extraction based on the selection criteria. Any potential disagreement was discussed comprehensively to obtain a final consensus. The main features of included studies are summarized, which includes: first author’s name, year of publication, origin and race, type of cancer, age range, total number of participants, genotyping assay of MMP-8 C-799 T, Lys460Thr, and Lys87Glu polymorphisms and cancer susceptibility. A total of five genetic models were adopted: allelic contrast , homozygote model , heterozygote model , dominant comparison , recessive comparison . Q-test was utilized to estimate the heterogeneity among enrolled researches. If the heterogeneity was absent (P > 0.05), the fixed-effects model was employed [P values of Begg’s and Egger’s test more than 0.05 indicated the absence of publication bias. The STATA software  was adopted for all the above analyses.We calculated the OR with 95% CI to investigate the strength of the relationship between employed ; alternaemployed . StratifMMP-8 has an impact on tumorigenesis, we employed the online TCGA database to evaluate the MMP-8 expression in bladder cancer tissue and control counterparts. The effect of the expression of MMP-8 on bladder cancer patients’ overall survival time was also assessed. Furthermore, we adopted bioinformatics tools, like Polyphen-2 (http://genetics.bwh.harvard.edu/pph2/), to predict the role of MMP-8 SNPs at the protein level.To further investigate whether the expression of MMP-8 C-799 T, Lys460Thr, and Lys87Glu variants have been summarized in Table MMP-8 polymorphisms in compliance with the inclusion criteria were finally identified in the present analysis. All eligible studies had NOS score more than 6. There were 4372 cancer patients and 5066 control participants for the analysis of MMP-8 C-799 T variant. Five articles were acquired for assessing the association of MMP-8 Lys460Thr variant on cancer susceptibility, including 3019 cases and 3544 control subjects. There were 749 cases and 1919 controls on the Lys87Glu polymorphism. The MAFs  of MMP-8 C-799 T variants were shown in Fig. MMP-8 Lys460Thr polymorphism: African, 0.222; East Asian, 0.009; European, 0.044; South Asian, 0.060; and American, 0.060. The MAFs for Lys87Glu variant were: African, 0.290; East Asian, 0.432; European, 0.453; South Asian, 0.400; and American, 0.480. In stratified analysis by race, 13 case-control studies were conducted on Asian descendants; five were based on the Caucasian population and one was based on Latin descendants. In stratified analysis by cancer type, four studies concerned bladder cancer. The rest were focused on other cancers, such as lung cancer, hepatocellular carcinoma, malignant melanoma, oral cancer, ovary cancer, and gastric cancer. In stratified analysis by the source of control, 12 studies were population-based controls, and the rest seven were hospital-based studies.Main characteristics of the eligible articles as well as the genotyping assay results of MMP-8 polymorphisms and cancer risk are provided in Table MMP-8 variants and cancer susceptibility was not significant. The MMP-8 C-799 T variant is not associated with the susceptibility of cancer under all genetic models . In subgroup analysis by cancer type, we also indicated no relationship between MMP-8 C-799 T variant and bladder cancer  or other cancers . In stratified analysis by race and source of control, no significant association between this polymorphism and cancer susceptibility was demonstrated ; homozygote model ; heterozygote comparison ; dominant model , and recessive comparison .Summarized results and details of the present analyses for the three MMP-8 expression was elevated in bladder cancer tissue as compared to their control counterpart , and Lys87Glu  variants and protein damage. Mutations of these SNPs are predicted to be “BENIGN” with a score less than 0.05, which indicated that neither Lys460Thr nor Lys87Glu SNP may probably damage the protein of MMP-8 . For MMP-8 Lys460Thr variant: M-allele vs. W-allele, t = − 0.35, P = 0.752; MW vs. WW, t = − 0.71, P = 0.527; MM + MW vs. WW, t = − 0.68, P = 0.545. For Lys87Glu variant: A vs. G, t = 0.38, P = 0.771; AA vs. GG, t < − 0.01, P = 0.998; AG vs. GG, t = − 0.04, P = 0.975; AA + AG vs. GG, t = − 0.13, P = 0.916; AA vs. AG + GG, t = 1.71, P = 0.338. Outlines of the funnel plots were relatively symmetrical for overall cancer risk, implying no significant publication bias . Our finding was in agreement with the studies conducted by Hsiao et al., Huang et al., and Wieczorek et al [Pheterogeneity = 0.286, P = 0.671) and other cancers . In stratified analysis by race, a similar result was indicated in Asian  and Caucasian descendants . For MMP-8 Lys460Thr variant, no positive correlation was found in the overall analysis . Similar results were indicated for Lys87Glu variant . Results from in silico analysis showed that MMP-8 expression was elevated in bladder cancer tissue as compared to the control counterpart. However, both the higher and lower MMP-8 expression groups did not have an impact on the patients’ overall survival time. Moreover, Polyphen-2 bioinformatics tool was also adopted to confirm the results of our present analysis. As the report for MMP-8 C-799 T (rs11225395) variation was not available, the association between Lys460Thr , and Lys87Glu  variants and protein damaging was further investigated. Mutations of Lys460Thr and Lys87Glu were predicted to be “BENIGN” with a score less than 0.05, which indicated that these SNPs do not damage MMP-8 protein, and are in agreement with the conclusions of the current analyses.In the present study, a total of 8140 patients with malignant carcinoma and 10,529 control participants were investigated. For ek et al , 31, 34.MMP-8 Lys460Thr SNPs and three for Lys87Glu polymorphism were acquired based on the selection criteria. As regard to C-799 T variant, only two case-control studies were focused on Caucasian population and one was based on Latin descendants. In addition, tumor stage and grade may potentially influence the results of the present analysis. We tried to further evaluate this effect in more details; however, raw data of eligible studies remains insufficient. More efficient investigations are still required to further strengthen the statistical power. Last but not least, P value for HWE was less than 0.05 in five of the included articles [In addition, several limitations of the present study should be clarified. First of all, the number of enrolled studies for subgroup analysis remains insufficient, which exhibits fairly limited statistical power. Only five studies for articles , 33–35, Despite these limitations, some key advantages should be acknowledged. First, all eligible case-control studies according to the selection criteria were obtained and the statistical efficiency was enhanced remarkably. Second, no obvious publication bias was indicated by Egger’s and Begg’s funnel plot, which showed that the findings of the current analysis can be considered reliable. Additionally, NOS scores of the enrolled studies were more than 6, which indicated a high methodological quality of each article.MMP-8 C-799 T, Lys460Thr, and Lys87Glu variants are not participant with the susceptibility of cancer. Further well-designed investigations are still warranted to confirm this conclusion in more detail.Taken together, based on the currently published data, our study showed evidence that


Furthermore, we showed a different frequency of the repeat between the northern and southern European populations . Our findings provide evidence for the association between intermediate repeats and ALS  with direct relevance for research and clinical practice by showing that an expansion of 24 or more repeats should be considered pathogenic.The expansion of a hexanucleotide repeat GGGGCC in C9orf72 is the most common known cause of ALS accounting for ~ 40% familial cases and ~ 7% sporadic cases in the European population. In most people, the repeat length is 2, but in people with ALS, hundreds to thousands of repeats may be observed. A small proportion of people have an intermediate expansion, of the order of 20 to 30 repeats in size, and it remains unknown whether intermediate expansions confer risk of ALS in the same way that massive expansions do. We investigated the association of this intermediate repeat with ALS by performing a meta-analysis of four previously published studies and a new British/Alzheimer’s Disease Neuroimaging Initiative dataset of 1295 cases and 613 controls. The final dataset comprised 5071 cases and 3747 controls. Our meta-analysis showed association between ALS and intermediate C9orf72 repeats of 24 to 30 repeats in size  contains supplementary material, which is available to authorized users. Amyotrophic lateral sclerosis (ALS) is a neurodegenerative disease, primarily affecting upper and lower motor neurons, resulting in progressive weakness and culminating in death from neuromuscular respiratory failure, typically 2–5 years after diagnosis . The incC9orf72 gene was found to be the most frequent cause of ALS [Following initial linkage and association studies, the massive expansion of a hexanucleotide repeat in the e of ALS –10. In me of ALS . A smalle of ALS –16. Howee of ALS .p-value = 0.03, fixed-effect model) [Q-test p-value = 0.61), heterogeneity may be masked. Genetic effects may vary across different populations for various reasons, including both genuine differences and differential biases and errors across studies [C9orf72 repeat expansion exemplifies this variation [Q-test and I-squared do not describe heterogeneity well when the number of studies is small [One meta-analysis based on four studies, suggested an association of intermediate expansions between 24 and 30 repeats in size with sporadic ALS (chi-squared t model) . Howevert model) . Second,t model) , 19. Whi studies , 21, andariation , 22, 23.is small  and the is small , 7, 14.C9orf72 repeats, and included the findings in a meta-analysis of studies for which there are data on intermediate repeats of size 24 or greater. We did not investigate intermediate repeats of size 20 to 23 since such repeats have been observed both in ALS patients and controls with no differences in allele distribution in several studies [Table S1).Taking these considerations into account, we therefore analysed a new group of 1295 people with ALS and 613 controls, all sized for  studies , 23, inchttp://adni.loni.usc.edu). The UK dataset was generated as part of Project MinE [Whole-genome sequencing (WGS) data were obtained from blood samples of 1908 individuals, comprising 1295 people with ALS  and 613 unaffected controls from two groups: a UK set of 1295 ALS cases, 340 matched controls , and 273ect MinE , 27, on C9orf72 GGGGCC repeat in the WGS data. ExpansionHunter has been previously validated for C9orf72 repeat sizing in 5787 WGS samples from Project MinE that were also genotyped using repeat-primed polymerase chain reaction (PCR) and fluorescence PCR. Using the repeat-primed PCR calls as the gold standard, ExpansionHunter showed an expansion detection accuracy > 99% [We used ExpansionHunter  to size cy > 99% . Expansicy > 99% , 29, whiI-squared statistic and chi-square-based Q-tests. The Q-test power was calculated using the Jackson method [Q-test power calculation was made using Stata 15 . All other analyses were done in R 3.4.4 using the Metabin function of the Meta library [p-value threshold used in this study to indicate statistical significance was 0.05.For meta-analysis, we selected studies for which data were available for repeats of size 24 or more . Associan method . Both fi library . The p-vData used in the preparation of this article were obtained from the Alzheimer’s Disease Neuroimaging Initiative (ADNI) database (adni.loni.usc.edu). The ADNI was launched in 2003 as a public-private partnership, led by Principal Investigator Michael W. Weiner, MD. The primary goal of ADNI has been to test whether serial magnetic resonance imaging (MRI), positron emission tomography (PET), other biological markers, and clinical and neuropsychological assessment can be combined to measure the progression of mild cognitive impairment (MCI) and early Alzheimer’s disease (AD).For the British dataset, the median age of ALS onset was 62.4 years and for controls, 60.1 years, with a male-female ratio of 62:38. In the ADNI dataset the median age was 73.2 and the male-female ratio 49:51.p-value = 0.96). Nine cases and one control were found to carry 24–30 repeats. Five controls were found to carry expansions whose length is rarely seen in unaffected controls and among which a few were hundreds of repeats long . Specifically, five British controls carried large expansions up to hundreds of repeats  and one British control carried an intermediate repeat expansion of 26 repeats. 6.6% of cases had a repeat of size > 30.We identified no difference in the distributions of repeat size between ALS and controls in the 1–23 repeat interval    , but this was underpowered to detect heterogeneity (power = 12.07%), suggesting that a random-effects model is a more appropriate model to use. Under a random-effects model, we did not show association using the four studies from the previous meta-analysis only . Using the UK/ADNI dataset we were able to show this association under the random-effects model hypothesis also   or Begg  test was observed . Sensitivity testing, recalculating the pooled odds ratios by omitting one study per iteration  showed that results were consistent, without major fluctuations.No evidence of publication bias from either the Egger  and the Fisher’s exact test . Considering that the frequency of the C9orf72 expanded repeat greatly varies across European populations [p-value = 5 × 10− 3, odds ratio = 6.17, 95% confidence interval 1.43–26.61) and the Fisher’s exact test  showed a significant difference between the two groups. On the basis of intermediate repeat heterogeneity between the two groups, we tested the association between the intermediate repeats and ALS in the two groups separately. For the northern European dataset (3015 cases and 2637 controls) the resulting odds ratio was 15.83 . In the southern European dataset (2056 cases and 1110 controls) the association was not significant . However, the low frequency of intermediate expansions (< 0.1% in cases and not present in controls) limits the statistical power to conclude if there is an association in this group (power < 10%).We also investigated the association between the intermediate repeats and ALS in a final dataset generated from the aggregation of the 5 datasets used for the meta-analysis (5071 cases and 3747 controls). The association was demonstrated both with the chi-squared test  and a younger age at onset  compared to patients carrying a non-expanded number of repeats . No difference in terms of gender prevalence and site of onset were observed between the three groups. A demographic table for the three groups is available in the Supplementary material .For our 1295 British patients, we compared the age at onset, time from first symptom to death (or last visit for censored cases) and site of onset , between the patients who carried a non-expanded number of repeats (< 24 in 1201 patients), an intermediate number of repeats (> 23 and < = 30 in 9 patients) and a full expansion (> 30 in 85 patients). We found no significant differences between the intermediate repeat carriers and the other patients. Consistent with previous findings , 32–34, C9orf72 gene are associated with ALS. An expanded hexanucleotide repeat in C9orf72 is the most common cause of ALS. While expansions of hundreds of repeats in size are clearly associated with the disease, a lower limit has not yet been identified. A few studies have suggested that 24–30 repeats might be sufficient to cause the disease [p-value = 0.01) and a random-effects model . The latter tells us that the association between 24 and 30 repeats and the disease exists without assuming that the effect of this mutation is homogenous across the patient groups. This up-to-date figure obtained from a meta-analysis on 5071 ALS patients and 3747 controls from 5 populations  provides definitive evidence for the association between intermediate repeats and ALS  and expands the role of C9orf72 in the development of ALS by enlarging the number of cases that can be explained, with direct implications for research and clinical practice. Indeed, we suggest the threshold to discriminate between neutral and pathogenic C9orf72 expansions should be lowered to 23, considering as pathogenic, expansions of 24 or more repeats, in agreement with previously reported observations [We have shown that intermediate repeat expansions in the  disease , 17. HowC9orf72 repeat) of over 10,000 thousand people with ALS, will be crucial to this end. Interestingly, we also observed a few controls with very large expansions that are only rarely observed in non-affected individuals. We have openly released the repeat frequency in our cohort for a wide range of lengths, to allow further analysis when new cohorts become available . Finally, we looked at age at onset, survival time, gender prevalence and site of onset of the intermediate repeat carriers, and we found no significant differences from the other patients. However, this analysis was limited by their small number (9), which does not allow us to make any conclusion about the clinical characterization of this sub-group of patients.Expansions of 20–23 repeats have contradictory evidence of association with ALS. This suggests that if such an association exists, their contribution to disease risk is weaker than for longer expansions, and we need larger datasets to investigate it. Due to the very low frequency of intermediate repeats, further investigation and larger sample sizes are needed to consolidate our results and determine whether to lower the threshold further. International initiatives such as Project MinE, whose aim is to collect genetic data  Egger test; B) Begg test. C) Funnel plot of the 5 studies used in our meta analysis. Table S1.C9orf72 expansion analysis results obtained with ExpansionHunter on the British/ADNI dataset. Table S2. Pooled odds ratios by omitting one study per iteration. Demographic table. (DOCX 3047 kb)


MTHFR polymorphisms with the clinical response to fluoropyrimidine-based chemotherapy has been explored, but the results remain controversial. Thus, a meta-analysis was performed to provide a comprehensive estimate in this account. Relevant studies were identified through PubMed, Embase and Web of Science databases from inception up to May 2017. Odds ratios (ORs) with corresponding 95% confidence intervals (CIs) were applied to assess the strength of association. A total of 2118 colorectal cancer patients from 21 studies were included in the meta-analysis. Overall, there was no significant association between MTHFR C677T (rs1801133) or A1298C (rs1801131) polymorphisms and the clinical response to fluoropyrimidine-based chemotherapy under all of the three genetic models  and stratification analysis, except for the retrospective study subgroup in the dominant model of MTHFR C677T and the “5-Fu + FA” treatment group in the allele contrast of MTHFR A1298C. No or moderate heterogeneity was observed in all genetic models. This meta-analysis suggested that MTHFR polymorphisms could not be considered as reliable factors for predicting the clinical response to fluoropyrimidine-based chemotherapy in colorectal cancer patients.Methylenetetrahydrofolate reductase (MTHFR) is a critical enzyme influencing the metabolism of fluoropyrimidines. The relevance of  Fluoropyrimidines, such as 5-fluorouracil (5-FU), and the oral prodrugs capecitabine and tegafur have been widely used in the treatment of a variety of solid cancers for a long time, especially colorectal cancer (CRC) –3. FluorMTHFR locates in chromosome 1p36.3, and is highly polymorphic [MTHFR gene, C677T  and A1298C , have been identified the main variants affecting the activity of this enzyme [MTHFR C677T and A1298C polymorphisms are potential predictors for the clinical response to fluoropyrimidine-based chemotherapy.Methylenetetrahydrofolate reductase (MTHFR), the most critical enzyme in folate-metabolizing pathway, catalyzes the irreversible conversion of 5, 10- methylenetetrahydrofolate  to 5-methyltetrahydrofolate, and reduces the amount of 5, 10-MTHF available for binding to FdUMP and TS [MTHFR gene polymorphisms are closely related to the efficacy of fluoropyrimidines, in fact the available evidence from the genetic association studies in clinic was weak and the published results were inconsistent. This discordance also existed in the existing meta-analysis. The systematic review conducted by Elias Zintzaras and colleagues indicated that MTHFR C677T and A1298C gene polymorphisms could not be considered as reliable predictors of response to fluorouracil-based chemotherapy in patients with colorectal cancer [Although theoretically l cancer , 11. In MTHFR C677T or A1298C polymorphisms and response to fluoropyrimidine-based chemotherapy in CRC patients were collected for the meta-analysis [MTHFR C677T polymorphism, and 13 studies provided 1496 patients for testing the association of MTHFR A1298C variant with response to chemotherapy . The flowchart of study selection was shown in Figure analysis . The chaMTHFR C677T were summarized in Table MTHFR C677T polymorphism and response to fluoropyrimidine-based chemotherapy under all three genetic models: allele model  , and no significant heterogeneity under recessive model . When stratified by ethnicity, study type, clinical outcome and chemotherapy regimen, only the retrospective study subgroup in dominant model showed a significant association . Moreover, the heterogeneity only reduced simultaneously when stratified by ethnicity under allele model  Figure , dominan) Figure , and rec) Figure . The Q-sMTHFR A1298C polymorphism and response to fluoropyrimidine-based chemotherapy, the pooled results from meta-analysis showed no significant association in all of the three genetic models , with fluctuation in the period of 2006–2012. The significant association just existed before 2005  was positively linked to response, with response rates of 37.1%, 53.3%, 62.5% and 80.0% in CRC patients bearing no, one, two or three favourable alleles, respectively [MTHFR and the efficacy of fluoropyrimidine-based treatment [in vitro studies [MTHFR C677T and A1298C polymorphisms on fluoropyrimidine-based chemotherapy in CRC patients, a meta-analysis including 21 studies was performed. The results of present meta-analysis suggested that there was no significant association between MTHFR C677T or A1298C polymorphisms and the clinical response to fluoropyrimidine-based chemotherapy in patients with CRC under allele, dominant and recessive models. Of note, when stratified by study type and chemotherapy regimen, the significant association could be observed in the retrospective study group of MTHFR C677T dominant model and the “5-Fu + FA” treatment group of MTHFR A1298C allele contrast. However, taking into account the authority of retrospective studies and the small number of studies included in the analysis, the results of these subgroup analysis require further assessed more scientifically.The association of ectively . While treatment , 31. Sim studies , 34. To MTHFR C677T  studies involving any type of colorectal cancer; (b) using chemotherapy regimens containing 5-Fu, capecitabine or tegafur; (c) using validated molecular methods for genotyping; (d) providing information on MTHFR and genotyping methods. Any discrepancies in data extraction were resolved by consensus.Full reports of relevant studies were retrieved and independently extracted by two investigators (Yuan Zhang and Jun-Lan Chuan). The extracted data included first author's name, publication year, ethnicity of the study population, study design, distribution of gender and age in patients, clinical outcomes investigated, chemotherapy regimen, clinical response, genotype distribution of The quality of the included studies was evaluated independently by two reviewers according to the Newcastle-Ottawa Scale (NOS) . The NOSMTHFR C677T or A1298C polymorphisms and clinical response was assessed by odds ratio (OR) and corresponding 95% confidence interval (CI) under the allele model , dominant model , and recessive model . The heterogeneity between included studies was evaluated by the Q-test. P > 0.1 indicates that there is no significant heterogeneity. I2 statistic was also calculated to quantify the heterogeneity: I2 < 25 %, I2 = 25–50%, I2 = 50–75% and I2 > 75%, indicated no heterogeneity, moderate heterogeneity, large heterogeneity, and extreme heterogeneity, respectively. When QP > 0.1 and I2 < 25%, the heterogeneity was considered to be nonsignificant and then the pooled OR and 95% CIs could be assessed by the fixed-effects model; otherwise, the random-effects model was used. Subgroup analyses were performed based on ethnicity (Caucasians and Asians), study type (prospective and retrospective), clinical outcome (objective response and TRG) and chemotherapy regimen, and only for groups reported in at least three independent studies. The sensitivity analysis was carried out by sequential omission of individual studies to assess the stability of the results. The publication bias was detected using Egger's regression test and Begg–Mazumdar adjusted rank correlation test. P < 0.05 indicated the presence of potential publication bias. Additionally, the cumulative meta-analysis was also carried out chronologically by publication year to observe the trend in estimated risk effect. All statistical analyses were conducted with the software STATA version 12.0 .The strength of association between the