There was also a significantly negative association between high level of UCA1 and poor grade cancer  and positive lymphatic metastasis . In conclusion, our study suggested that UCA1 was correlated with more advanced clinicopathological features and poor prognosis as a novel predictive biomarker of patients with various tumors.Urothelial cancer associated 1 (UCA1) as an oncogenic long non-coding RNA (LncRNA) was aberrantly upregulated in various solid tumors. Numerous studies have demonstrated overexpression of UCA1 is an unfavorable prognostic indicator in cancer patients. This study aimed to further explore the prognosis role and clinical significance of UCA1 in cancer. Eligible studies were recruited by a systematic search in PubMed, Embase, Cochrane Library and Web of Science databases. A total of 19/16 studies with 1587/1291 cancer patients were included to evaluate the association between UCA1 expression and overall survival (OS) and clinicopathological factors of malignancies by computing hazard ratio (HR), odds ratios (OR) and confidence interval (CI). The meta-analysis indicated overexpression of UCA1 was significantly correlated with unexpected OS in patients with cancer (pooled HR = 1.85, 95% CI 1.62–2.10,  Non-coding RNAs (ncRNAs) encoded by the eukaryotic genome are considered as a large number of RNAs which are not translated into proteins. Indeed, more than 90% of noncoding RNAs used to be recognized as “biological noises” in transcription progression compared with about 3% protein-coding genes until the development of high-throughput sequencing technology and large-scale mapping of transcriptomes , 2. BaseUrothelial cancer associated 1 also known as UCA1 is located on 19p13.12 encodes 3 isoforms  with ployA tails, in which the 2.2 kb isoform is called cancer upregulated drug resistant (CUDR). Previous study has shown UCA1, a highly conserved nuclear-enriched lncRNA, is ubiquitous in different development stages of urinary system, productive system, digestive system and respiratory system . Cheng eFurthermore, increasing interest has been focused on whether UCA1 acts as a diagnosis and prognosis biomarker in cancer detection and treatment. UCA1 was firstly characterized as an effective diagnosis biomarker in bladder cancer with a high sensitivity and specificity (80.9% and 91.8%) . MeanwhiUpon an electrical search on PubMed, Embase, Cochrane Library and Web of Science databases, a total of 19 eligible OS papers including 1587 tissue specimens from 1774 records published from 2014 to 2017 were enrolled by a cautious searching strategy and full-text screening, which were based on inclusion and exclusion criteria in this meta-analysis Figure . The maip < 0.001) and no obvious heterogeneity existed across 19 studies under a fixed effect model  in Figure To assess the association between the UCA1 expression and OS of all cancer patients, a total of 1587 patients with HRs and 95% CIs were included. A significant association was observed between high UCA1 level and poor OS in patients with all 10 types of cancer , CRC , HCC , and PAC . There showed no statistical significance of heterogeneity test in subgroups of GC , CRC , HCC , and PAC . Then, a significant effect of elevated UCA1 on OS emerged in multivariable analysis subgroup  and univariable analysis subgroup . No statistically significant heterogeneity was in multivariable analysis subgroup  and univariable analysis subgroup . Meanwhile, we also detected a significant correlation between overexpression of UCA1 and poor prognosis of cancer patients in large specimen size subgroup  and small specimen size subgroup  with no evident heterogeneity  respectively. To the quality score subgroups, we obtained similar association of upregulation of UCA1 with OS in high quality score subgroup  and low quality score subgroup . No evidence of significant heterogeneity was found across 2 subgroups . Quantified heterogeneous test by meta-regression indicated that no specific factor accounted for the heterogeneity among studies of interest consistent with the outcomes of subgroup analysis in cancer type, analysis method, tumor sample size and quality score covariates.The subgroup analysis in a fixed effect model, random effect model and meta-regression was conducted by cancer types, analysis methods, sample sizes and quality scores for a subsequent investigation of potential heterogeneity in Table Egger’s = 0.006, Figure Egger’s = 0.001), high quality score (pEgger’s < 0.001) and multivariate analysis subgroup (pEgger’s < 0.001) in  in . Howeverp < 0.001), positive lymphatic metastasis , and distant metastasis  in Table In the clinicopathological studies, 16 researches consisting of 1291 tumor samples with a correlation between clinicopathological features and UCA1 expression were retrieved in OR analysis. Clinical stage, lymph node metastasis, distant metastasis and tumor size data were collected to analyze. UCA1 expression was significantly different in these clinicopathological factors . We foun2 = 56.5%, pH = 0.024) in 8 clinicopathological studies shown in Table 2 = 71.9%, pH = 0.014) subgroup. No evidence of statistical heterogeneity was detected in cancer type and lymphatic metastasis subgroup . However, only CRC showed clinical significance and no obvious heterogeneity  by stratification analysis of cancer type. In subsequent stratification analysis of lymphatic metastasis, there was no evidence of significant heterogeneity in CRC  and EOC  subgroup.The subgroup analysis stratified by cancer type, sample size and quality score uncovered the resource of heterogeneity under the fixed effect model, random effect model and meta-regression analysis Table , 4, 5, 6Egger’s = 0.622 and pEgger’s = 0.653, respectively) and no asymmetry among 4 subgroups in the sensitivity analysis . Meanwhile, the pooled data of eligible studies also indicated that high UCA1 expression was significant correlated with poor grade cancer  and positive lymph node metastasis .Here, we performed this meta-analysis to explore the correlation between high expression of UCA1 and clinicopathological characteristics and evaluate the prognosis role of UCA1 for cancer patients. All of these results above suggest that high UCA1 expression may be regarded as an unfavorable predictor in different cancers  with the following searching strategy: “UCA1” OR “UCA 1” OR “UCA-1” OR “urothelial cancer associated 1” OR “urothelial cancer associated-1” OR “CUDR” OR “LINC00178” OR “NCRNA00178” OR “onco-lncRNA-36” OR “ENSG00000214049”) AND (“cancer” OR “tumor” OR “tumour” OR “carcinoma” OR “neoplasm” OR “adenoma” OR “sarcoma”). We manually searched retrieved references for potentially missing literatures. The cited articles were excluded from duplicated firstly, then titles and abstracts were carefully scanned to eliminate irrelevant studies. Finally, we prudently reviewed full texts of potential retrieved studies. Studies were available upon the eligibility criteria: (1) showed the relationship between the dichotomous UCA1 levels and prognosis and clinicopathological characteristics of patients with any type of cancer; (2) calculated HRs and 95% CIs for overall survival (OS), progression-free survival (PFS), disease-free survival (DFS); (3) were published in English. Duplicated, non-dichotomous expression of UCA1 or non-English articles, absence of key survival outcome such as HRs, 95% CIs or Kaplan-Meier survival curve, reviews, letters, laboratory studies of non-human research or comments were omitted.Eligible data were deliberatively judged and double checked from available studies based on criteria of inclusion and exclusion. To each study, we carefully extracted the following information: first author, journal, year of publication, country, ethnicity of the study population, type of specimen, carcinoma type, number of patients, detection method, cut-off value, follow-up, quality score and HR and 95% CI for OS, PFS, or DFS. Quality assessment of the available studies was performed upon the REMARK guideline . HR, as Q test and Higgins I-squared statistic (PH < 0.1 and I2 > 50%). To further explore the potential heterogeneity factors and outcome stability in studies, a subgroup analysis and a sensitivity analysis were utilized respectively in the meta-analysis. Publication bias was evaluated by a funnel plot and Egger's linear regression test with a significant publication bias by P < 0.05.All analysis was conducted with STATA software version 12.0 . Odds Ratios (ORs) were performed to analyze the association of UCA1 expression with clinical stage, tumor size, lymphatic and distant metastasis and HRs with corresponding 95% CIs were utilized to estimate the relationship strength between UCA1 expression and patients’ prognosis. If the HRs were not directly reported in original articles, we calculated the essential data upon the previously reported methods . The poo


P-nonlinearity =0.031), suggesting that per 5 kg/m2 increment on BMI corresponded to a 3.1 % increase of bladder cancer risk, especially BMI exceed 30kg/m2. Furthermore, significant positive association was also observed between obesity category and bladder cancer risk . In summary, this dose-response meta-analysis suggests a nonlinear positive association between BMI and bladder cancer risk. Further studies are required to confirm these findings and elucidate the pathogenic mechanisms.Prospective epidemiologic studies on the association between body mass index (BMI) and bladder cancer yielded inconsistent findings. This study sought to quantitatively summarize the evidence by performing a dose-response meta-analysis on prospective cohort studies. Eligible studies were retrieved via PubMed and Embase databases, and by manual review of the references. Linear and nonlinear trend analyses were conducted to explore the relationships between BMI and bladder cancer risk. Meta-analyses on the categories of overweight and obesity were also conducted. The summary relative risk (SRR) was estimated. Heterogeneity across the studies was explored through subgroup analyses based on gender, age, year of publication, sample size, assessment of BMI, geographic location, physical activity and family history of cancer. A total of 14 prospective cohort studies involving 12,642 cases were included. Result of the dose-response analysis showed a nonlinear positive relationship between BMI and bladder cancer (SRR = 1.03, 95% CI: 1.01-1.06,  Bladder cancer is among the most common malignancies of the genitourinary tract all over the world . ApproxiExcess bodyweight, whether in people who are overweight or obese, is increasingly recognized as an important risk factor for many common cancers . Body maThus, we conducted this meta-analysis to quantitatively summarize the evidence on the association of BMI with bladder cancer risk by performing a dose-response meta-analysis on the basis of prospective cohort studies.via database search. Of these, 2453 articles were excluded considering of potential value after excluding the duplicate or not relevant articles, mechanism studies, case reports or reviews. Furthermore, 106 articles were excluded following an initial screen of abstracts, and the other 11 papers were excluded after reviewing full-texts because of a lack of results reported BMI and bladder cancer. The remaining 9 articles finally met all the selection criteria. Of note, 5 additional articles were included from references review. Thus, a total of 14 articles  with a sample size of 5,640,760 participants were included in our meta-analysis.Figure The main characteristics of the included studies are shown in Table 2 increase in BMI for each studies, the dose-response meta-analysis in random-effects model across all the studies showed that there was a weak positive association between BMI and bladder cancer risk, indicating that per 5 kg/ m2 increment on BMI corresponded to a 3.1 % increase of bladder cancer risk (The summary relative risk (SRR): 1.03, 95%CI: 1.01-1.06, p = 0.039) with evidence of no problems in the fitted model . There was evidence of a potential nonlinear association between BMI and risk of bladder cancer, which showed that there seem to be a slight higher risk in the range of 24-28 kg/m2 and a further, with the curve becoming steeper and almost linear after the BMI exceed 30kg/m2 compared to the average the reference of 22kg/m2 , but not for overweight category . These results indicated that overweight was not associated with bladder cancer risk, but obesity was associated with increased by 10.0% risk of bladder cancer , and obesity , respectively.Compared to normal weight, meta-analyses based on BMI categories found that the SRR was 1.03 (95%CI: 0.95-1.11) for overweight, and 1.10 (95%CI: 1.03-1.17) for obesity. There was evidence of heterogeneity among studies for obesity category . We performed sensitivity analyses and found that none of the studies considerably affected on the overall or stratified estimate between BMI and bladder cancer in this meta-analysis (data not shown). In addition, little variability was observed among the most subgroup analyses.It has been reported that BMI has been associated with worse outcomes in several solid malignancies. Weight gaining is known to predispose to a number of cancers in certain populations, including cancer of colon, kidney, breast, endometrium, gallbladder and so on . However2 increase. Further our finds indicated that a positive association was found between bladder cancer and obesity compared with normal weight, with little variability. Considering the unsatisfactory exploring in single study and only a weak association between BMI and bladder cancer risk was detected based on linear trend analysis, we conducted a nonlinear trend analysis which also showed a significant fitting, and implying that further meta-analyses based on categories were necessary. The results of meta-analyses based on overweight and obesity confirmed our assumption that bladder cancer risk was increased in particular range of BMI. Notably, no significant association was observed between overweight category and bladder riskIn this updated meta-analysis, fourteen prospective cohort studies comprising 5,640,760 participants were included in the current meta-analysis to explore the dose-response relationship between BMI and bladder cancer risk. Overall, the dose-response meta-analysis in random-effects model across all the studies showed that there was a potential nonlinear association between BMI and bladder cancer risk and the risk increased by 3.1% for each 5 kg/mA greater heterogeneity was detected in those studies on obesity category likely due to differences among study populations, model selection, analytic methodology and exposure assessment. Subgroup analyses were conducted to investigate the sources of the observed heterogeneity among studies. The results in subgroup analyses were consistent with the overall results. A significant proportion of the observed heterogeneity may be explained through the subgroup analyses based on gender, age, year of publication, sample size, assessment of BMI, geographic location, physical activity and family history of cancer. Little variability was observed through the stratified analyses. Furthermore, we have, however, shown an absence of publication bias in these meta-analyses with either the Egger’s or Begg’s tests. Sensitivity analyses showed none of the studies considerably affected the summary associations between BMI and risk of bladder cancer.Several biologic mechanisms contribute to the association between excess weight and risk of cancers, including insulin resistance, resultant chronic hyperinsulinaemia, increased bioavailability of steroid hormones and localized inflammation . The undThe present meta-analysis has several key strengths. The present meta-analysis was only based on well-established prospective cohorts which had minimized recall and selection biases. The included studies with large sample sizes and long term follow-up durations enhanced the statistical power to detect more stable associations and provide more reliable estimation. If a true association between overweight and bladder cancer risk existed, the follow-up durations in our analyses should have been long enough to detect such an association. Furthermore, we have examined the exact shape of the dose-response relationship between BMI and bladder cancer risk. Another strength is the robustness of the findings from the multiple subgroup analyses. The overall findings were consistent with the results of the subgroup analyses independent of gender, age, year of publication, sample size, assessment of BMI, geographic location, physical activity and family history of cancer.via subgroup analyses based on gender, age, year of publication, sample size, assessment of BMI, geographic location, physical activity and family history of cancer. Furthermore, our sensitivity test showed the findings were robust, and potential bias was not suggested by the Begg’s or the Egger’s test. Second, all the included studies in our analysis were prospective cohort studies. However, differences among follow-up duration, assessment of BMI and measurement of bladder cancer end point, may hinder an estimate of the true effects. Third, residual confounding factor is always a major concern in the epidemiology studies. The majority of included studies had adjusted with some lifestyle variables, but we could not exclude the possibility that other uncontrolled or unmeasured confounding factors play roles in the summary associations. An additional limitation is that the results of our meta-analysis have limited generalisation to other regions (e.g. Latin America and Africa), though the geographical regions covered in this meta-analysis included Asia Europe, the USA and some multi-international centers.Our study has some limitations when we interpreting the current results. First, although heterogeneity was observed across studies when we conducted meta-analysis, we were able to reveal the major sources of heterogeneity 2 increment on BMI, especially when BMI was higher than 30kg/m2. Given the alarm rising rates of overweight and obesity worldwide, further studies are required to confirm these findings and elucidate the pathogenic mechanisms.To sum up, our dose-response meta-analysis on the basis of prospective cohort studies indicated an increased bladder cancer risk of 3.1% for each 5 kg/mTwo authors independently and systematically searched Pubmed and Embase databases up to December 01, 2016. Research articles were selected using the following terms: ”body mass index”, or ”BMI ”, or ”obesity”, or ”adiposity”, or ”overweight”; “bladder cancer”, or ”urinary bladder neoplasms” or ”urothelial carcinoma”. The search was focused on human studies, without any other restriction. We also scrutinised the reference lists of reviews, meta-analyses, and selected research articles to identify additional relevant studies. Additionally, in view of the large number of bladder cancers arising in China, we have also searched the China National knowledge infrastructure (CNKI).We included studies that met the following criteria:(a) An original article; (b) A prospective cohort study; (c) Determining BMI at baseline and then recording the incidence of bladder cancer during follow-up; (d) Risk estimate as hazard ratio (HR), risk ratio (RR) or odd ratio (OR) with corresponding 95% confidence interval (CI) for more than three categories of BMI or providing sufficient data to estimate these. (e) RR and corresponding 95% CI at least with adjustment for age. If the studies were reported from the same or overlapping cohort, only the most recent and informative one would be included. Discrepancies between two investigators were solved by discussion.Two investigators independently extracted data, which were cross-checked by another investigator. A standardized data-collection protocol for each study included was used with following information: the first author, the year of publication, gender, study area, duration of follow-up, sample size, number of cases, ascertainment of exposure, cut off of exposure, and adjustments for confounders. If one study reported several risk estimates, we used the one from the main multivariable model which included more adjusted confounders. If the study didn’t provide total person-years or total participants for each group, we estimated them using the method carried out by Aune . In addi2), Overweight (BMI 25~30 kg /m2), and Obesity (BMI ~30 kg/m2) [BMI categories were defined according to the World Health Organization (WHO): Normal (BMI 18.5~25 kg/m0 kg/m2) .p value < 0.1 was considered statistically significant. Heterogeneity among studies was assessed using the Q or I2 statistics, which tested total variation across studies that was attributable to heterogeneity rather than to chance [Meta-analyses were conducted to explore the relative risk of bladder cancer on overweight and obesity compared with normal BMI level. For studies that reported RR for several categories of BMI that fell into the category representing overweight or obesity, we pooled the RR weight by the inverse of their variance. Furthermore, meta-analysis with the assumptions of a random-effects model was conducted to calculate The summary relative risk (SRR), which incorporates between-study variability into the calculation . A two-to chance .2kg/m increase on BMI by used of the method of generalized least-squares for trend estimates.9 The potential nonlinear dose-response relationship between BMI and bladder cancer risk was examined using a two-stage hierarchical regression model [A dose-response analysis was conducted based on the category data of BMI, number of cases, person-years and logarithm of RRs and its corresponding standard error. The eligible studies should provide sufficient information across at least three categories of exposure. Among the studies, we assigned a median of BMI for each category. For the open-ended upper category, the amplitude was assumed the same as the previous one. We transformed category-specific risk estimates into estimates of the relative risk (RR) associated with every 5 on model . Data weon model .Sources of heterogeneity were explored by subgroup analyses based on gender, age, year of publication, sample size, assessment of BMI, geographic location, physical activity and family history of cancer. Sensitivity analyses were conducted to clarify whether the results were affected due to one single study by repeating the meta-analysis after omitting one study at a time. We inspected the funnel plots for asymmetry and with Egger’s test and Begg’s test to test publication bias. All statistical analyses were performed with STATA Statistical Software, version 12.0.


However, the results remain inconclusive and controversial, particularly for the data derived from different ethnicities and genders. Therefore, we performed a meta-analysis to evaluate this association. All eligible case-control studies that met the search criteria were retrieved from multiple databases, and five case-control studies were included for detailed evaluation. The pooled odds ratios (ORs) with 95% confidence intervals (95% CIs) were calculated to assess the strengths of the associations of HTR3A (rs1062613) and HTR3E (rs62625044) polymorphisms with IBS-D risk. Our results revealed statistically significant associations of the HTR3A  polymorphism with a decreased risk of IBS-D in all genetic models. Additionally, the HTR3E  polymorphism was also found to be significantly associated with a decreased risk of IBS-D in the allele and recessive models. Subgroup analysis revealed that these associations held true especially for Asians and female. In conclusion, this meta-analysis suggested that the C allele of HTR3A (rs1062613) and the G allele of HTR3E (rs62625044) are associated with a decreased risk of IBS-D.Several studies have reported an association between serotonin receptor type 3 (5-HT Accoy IBS-D) –14.3 receptor is a Cys-loop ligand-gated ion channel composed of five subunits [HTR3A, HTR3B, HTR3C, HTR3D and HTR3E genes [HTR3 variants have been reported to be associated with schizophrenia [HTR3 polymorphisms (HTR3A and HTR3E) and the risk of IBS-D have been intensively investigated. However, the community is still unable to reach a consensus, particularly regarding the data from different ethnicities and genders [HTR3A and HTR3E polymorphisms on the risk of IBS-D.The 5-HTsubunits , which a3E genes –18. Diffophrenia , depressophrenia , anxietyophrenia , autism ophrenia , obsessiophrenia , drug abophrenia , 25. Rec genders –14, 26. HTR3A  polymorphism, 5 studies were available and included a total of 1,178 cases and 1,124 controls. Regarding the HTR3E  polymorphism, 5 studies involving a total of 1,276 cases and 1,389 controls were available. All of these included studies were hospital-based, of which 3 from China, 1 from Germany, 1 from USA. Additionally, all of the included studies were of high quality, as indicated by the Newcastle-Ottawa scale (NOS) scores of each study being above 6 points  and HTR3E  were estimated from the 1000 Genomes. The present samples indicated that 97.8% power for HTR3A  (MAF = 0.25), and 17.5% power for HTR3E  (MAF = 0.01). The power analysis indicated that these recruited samples could provide sufficient power in identifying the association between HTR3A  polymorphism and IBS-D.Before implementation of this meta-analysis, statistical power was assessed with the assumptions:α err prob = 0.05, OR = 1.30 (corresponding to a “weak to moderate” gene effect) for the two single nucleotide polymorphisms (SNPs), and minor allele frequencies (MAF) of HTR3A  and HTR3E  were used in our analysis.As illustrated in Figures P = 0.02; in the codominant model, CC vs. TT, OR = 0.36, 95% CI: 0.25–0.53, P < 0.00001; CT vs. TT, OR = 0.42, 95% CI: 0.28–0.64, and P < 0.0001; in the dominant model, CC+CT vs. TT, OR = 0.38, 95% CI: 0.26–0.56 and P < 0.0001; and recessive  , while no significant association was found among non-Asians. In addition, analysis stratified by different gender showed a significant association of IBS risk with the C allele deletion in both female and male. All the results are listed in Table The C allele of rs1062613 was found to be significantly associated with a decreased risk of IBS-D. The following data were obtained: in the allele model, C vs. T, OR = 0.73, 95% CI: 0.56–0.95, and P < 0.00001) and recessive model   and non-Asians . In addition, analysis stratified by different gender showed a significant association of IBS-D risk with the G allele deletion in female, but not in male. All the results are listed in Table The G allele of rs62625044 was found to be significantly associated with a decreased risk of ISB-D in allele model  and IBS-D. In Europe, Kapeller J [HTR3E variant c.*76G>A (rs62625044) is significantly associated with female IBS-D in two independent cohorts from the UK and Germany, while the HTR3A variant c.-42C>T (rs1062613) is only associated with the risk for IBS-D in UK, but not in Germany population. In Asia, Gu et al. [HTR3A and HTR3E are associated with the risk for D-IBS in Chinese population, especially in women. However, another studies [HTR3A polymorphism loci rs1062613 may be not associated with IBS risk in American population. Due to relatively small samples from different populations, these studies demonstrated inconsistent results. Therefore, we performed a meta-analysis to estimate the association between HTR3A and HTR3E polymorphisms and the risk for IBS-D. To the best of our knowledge, this is the first meta-analysis to explore the relationships between HTR3A and HTR3E gene polymorphisms and IBS-D susceptibility. Five case-control studies [HTR3A and HTR3E gene polymorphisms.Recently, several genetic association studies identified a novel association between peller J  found thT rs10626 is only  studies –14, 26 wHTR3A (rs1062613) decreases the risk of IBS-D in all of the comparison models, and the G allele of HTR3E (rs62625044) was also observed to have lower IBS-D risk in the allele and recessive models. Besides that, subgroup analyses by ethnicity indicated that SNP of rs62625044 (G/A) was significantly associated with a decreased risk of IBS-D in both Asian and non-Asian population, while the SNP of rs1062613(C/T) was only associated with the risk for IBS-D in Asian population. Moreover, analyses stratified by different gender showed a significant association between rs1062613(C/T) polymorphism and risk of IBS-D in both female and male population, while the SNP of rs62625044 (G/A) was only associated with risk of IBS-D in female population. Our results have some differences from the previous studies [In our meta-analysis, the main finding was that the C allele of  studies , 26. A pAdditionally, the genotype distributions in all of the controls were consistent with HWE, except one for one study reported by Kapeller et al. . HoweverThere were some limitations in the current study. Firstly, the sample sizes of the studies included in our meta-analysis were relatively small, especially in the subgroup of non-Asians, which may lead to a false negative result. Therefore, the negative results of the non-Asians should be interpreted with caution. Moreover, although meta-analysis has the benefit to overcome this limitation and may generate more precise results, the combined sample sizes in our meta-analysis were still inadequate to detect the association between rs62625044 polymorphism and IBS-D risk because power calculation for the pooled sample sizes was less than 80%, which might attenuate the statistical power to detect a slight effect and increase the chance of opportunity bias. Second, due to the limited data, we did not carry out subgroup analysis to other factors, which may participate in the progression of IBS-D, such as age, lifestyle and social psychology. Thirdly, only articles in English and Chinese language were included; thus, studies written in other languages were neglected. Finally, although we performed a systematic searching strategy to identify eligible studies, there was still probability that few studies so called “grey literatures” were not included.HTR3A (rs1062613) and the G allele of HTR3E (rs62625044) are associated with a decreased risk of IBS-D. One evidence indicates that these HTR3 polymorphisms can remarkably up-regulate the expression of 5-HT3 receptors, which have been proved to cause the gastric functional disorders including emesis, eating disorders and IBS-D, et al. [In conclusion, this meta-analysis suggested that the C allele of , et al. , as illuor “IBS”) AND  AND (“serotonin type 3” or “serotonin receptor 3” or “5-HT3” or “HTR3A” or “HTR3E”). Furthermore, the references of all retrieved articles were also checked by hand to identify additional potential studies. The languages were limited to English and Chinese.In accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement , we searHTR3A or HTR3E polymorphism and IBS; (2) cohort or case-control study (3) provided sufficient data of allele and genotype frequencies of SNPs or required information could be calculated; and (4) if serial studies on the same population were published, only the most recent study was included. Additionally, we excluded reviews, abstracts, and redundant and animal studies.The inclusion criteria were as follows: (1) studies of the association between ’s name, publication year, country of origin, ethnicity, sample size, genotyping method, and the HTR3A and HTR3E genotype distributions and alleles in the case and control groups.Two independent investigators extracted relevant data from all included studies on the basis of the inclusion criteria, and a third investigator verified them. The following information from eligible studies was extracted: the first authorThe quality of included studies were assessed by two investigators independently on the basis of Newcastle-Ottawa Scale (NOS) , which bP < 0.05 was considered statistically significant. Studies with the controls not in HWE were subjected to a sensitivity analysis [HTR3A and HTR3E polymorphisms with the risk of IBS were assessed by the pooled ORs with the corresponding 95% CIs under the following genetic models: allele model, codominant model, dominant model, and recessive model. The heterogeneity between studies was determined by the Cochrane’s Q-statistic test [I2 statistic. When I2 > 50% or PQ ≤ 0.1, which suggest substantial heterogeneity, a random-effects model (DerSimonian-Laird method) [P < 0.05 was considered statistically significant. All analyses were conducted using the RevMan 5.1 and STATA 12.0 software packages.The HWE of the genotype distributions in the controls of the include studies were tested by the Chi-square test, analysis . The powanalysis . The asstic test , and the method)  was used method)  was appl method) , and P <


Orofacial clefts include cleft lip (CL), cleft palate (CP), and cleft lip and palate (CLP), which combined represent the largest group of craniofacial malformations in humans with an overall prevalence of one per 1,000 live births. Each of these birth defects shows strong familial aggregation, suggesting a major genetic component to their etiology. Genetic studies of orofacial clefts extend back centuries, but it has proven difficult to define any single etiologic mechanism because many genes appear to influence risk. Both linkage and association studies have identified several genes influencing risk, but these differ across families and across populations. Genome-wide association studies have identified almost two dozen different genes achieving genome-wide significance, and there are broad classes of ‘causal genes’ for orofacial clefts: a few genes strongly associated with risk and possibly directly responsible for Mendelian syndromes which include orofacial clefts as a key phenotypic feature of the syndrome, and multiple genes with modest individual effects on risk but capable of disrupting normal craniofacial development under the right circumstances . Genomic sequencing studies are now underway which will no doubt reveal additional genes/regions where variants  can play a role in controlling risk to orofacial clefts. The real challenge to medicine and public health is twofold: to identify specific genes and other etiologic factors in families with affected members and then to devise effective interventions for these different biological mechanisms controlling risk to complex and heterogeneous birth defects such as orofacial clefts. Normally, development of the face and mouth occurs very early in pregnancy and reflects a complex process of cell growth and migration followed by fusion of symmetric structures to form the palate separating the mouth and nasal cavity, with the outer structures of the face developing before the inner structures. Collectively CL, CLP, and CP are the most common craniofacial birth defects worldwide, and affected individuals face feeding difficulties early in life and typically require multiple corrective surgeries, therapeutic dental procedures, and speech therapy throughout childhood4. In addition, individuals born with an orofacial cleft have increased incidence of mental health problems and higher overall mortality rates at all stages of life, even in developed countries with good medical care6. Where access to medical care is severely limited, infants born with CP or CLP generally have high mortality rates due to difficulty in breastfeeding, while untreated CL and CLP cases can face social discrimination throughout their lives. Thus, these birth defects have been subject to substantial selective pressure for most of human history.Orofacial clefts represent a group of anatomically distinct birth defects where there is a gap or break in normal features of the mouth, most commonly the roof of the mouth  or the upper lip or both. As shown in7. In general, East Asian and Native American populations have substantially higher birth prevalence rates than do European and South Asian populations, while African ancestry populations have lower birth prevalence rates10.Although clefts develop early in pregnancy, most epidemiologic data are based on cleft frequencies at birth. The worldwide average birth prevalence of all orofacial clefts is 9.92 per 10,000 , but there is substantial variation among populations12. Passive or “second-hand” exposure also appears to be a risk factor13. Evidence for maternal alcohol consumption as a risk factor is less consistent, although consuming large amounts over a short period of time (e.g. “binge drinking”) appears to increase risk14. Broad measures, e.g. socio-economic status, have been suggested as risk factors, but it is difficult to separate out the combined effects of maternal nutrition and health15. Several prescription medications have also been reported to increase risk when taken during the first trimester, including folate antagonists and certain other drugs exhibiting anti-folate properties7.There are several recognized environmental risk factors and multiple genes involved in the etiology of orofacial clefts, which means the etiology is complex and heterogeneous. Maternal smoking during pregnancy is a recognized risk factor for orofacial clefts, and it is estimated that 6.1%  of orofacial clefts could be avoided by eliminating maternal smoking16.Interestingly, a substantial number of studies suggest genetic and environmental risk factors may “interact” to modify risk to orofacial clefts. Several common exposures, including maternal smoking, alcohol consumption, and vitamin supplementation (a protective factor), have yielded statistical evidence of gene–environment (GxE) interaction, although the results from different studies are not always consistent. While biologically GxE interaction is eminently rational, replicating statistical evidence across studies is difficult and proving interaction exists remains a major challenge17 have recently reviewed the extensive history of genetic studies of orofacial clefts, which dates back to the 18th century  was 32 times higher than the baseline population risk, and for relatives of a CP case this relative risk was 56 times higher than baseline risk. This increased risk drops with increasing genetic distance: nationwide registry data from Denmark spanning the entire second half of the 20th century19 show the relative risk among all first-degree relatives  of CL/P cases was highest, while risks to second-degree relatives  were much smaller, and risks to third-degree relatives  were not significantly above the baseline population risk in Denmark (see their Table 2). When rules of exclusions and phenotype definition were adjusted between these population-based studies (Norway and Denmark), both gave similar patterns of strong familial aggregation, i.e. relatives of cleft cases always had a high relative risk compared to the population’s baseline risk, and there is a steep drop with increasing genetic distance between relatives.Orofacial clefts show strong familial aggregation, which suggests a major genetic component to their etiology. Marazita and Leslieet al.20 linked population-level vital records (birth certificate and twin registry data from Denmark) spanning most of the 20th century, achieving virtually complete ascertainment of cases for the latter part of the century. They showed MZ twins had much higher concordance rates than did DZ twins (47% MZ twins with 95% CI=31–64% compared to 8% for DZ twins with 95% CI=4–14%). These concordance rates yielded estimates of heritability (the proportion of variation in risk attributable to unobserved independent genes) of 91% for CL/P and 90% for CP. However, because the concordance rate among MZ twins is more than four times that of DZ twins, these Danish findings also raise the possibility that specific combinations of different genes (termed “epistasis”) may be important in the etiology of orofacial clefts.Twin studies of orofacial clefts consistently show higher concordance rates in monozygotic  twins compared to dizygotic  twins, again suggesting a major role for genes in controlling risk of being affected. Clearly, because MZ twins have identical genes over their entire genome, anything less than 100% concordance indicates some non-genetic component to the etiology of orofacial clefts, but whenever MZ twins consistently show greater concordance than DZ twins, it suggests some proportion of the variation in risk is under genetic control. Grosenth century by statisticians such as R.A. Fisher, W. Weinberg, and J.B.S. Haldane21. By the 1970s, segregation analysis had evolved to allow formal tests within a framework of “mixed models” that simultaneously considered straightforward Mendelian components (i.e. a “major gene”) and a more general residual “heritability” reflecting combined effects of many independent genes (collectively termed “polygenes”). A number of segregation analyses of orofacial clefts were performed during the early 1970s through the early 1990s for various racial/ethnic groups21. Collectively, however, this modeling approach rarely resulted in definitive and reproducible findings, and the most suitable model of inheritance remained the “multifactorial threshold model”, which is exquisitely vague in defining genetic mechanisms and statistically cannot discriminate between a single major gene vs. multiple “polygenes”. This multifactorial model hypothesizes a continuous “genetic liability” for risk, plus unspecified non-genetic risk factors that could also influence risk. Under this very general model, it is possible to estimate the “heritability” or the proportion of variation in risk attributable to any number of independent, autosomal genes, but it is not possible to estimate the penetrance or allele frequencies at any one of these separate risk genes.While clearly orofacial cleft risk shows strong familial aggregation, it does not strictly adhere to Mendel’s laws of inheritance, even within multiplex families (i.e. those with two or more affected members), which has bedeviled geneticists for decades. Statistical methods for fitting genetic models to family data are collectively termed “segregation analyses”, and these trace back to the foundations of human genetics. Statistical methods were initially developed in the early 2019. The sharp decline in relative risk to relatives of cases with increasingly distant relationships is also completely compatible with the multifactorial threshold model because the probability of sharing alleles identical by descent is constant whether one gene, a few genes, or many genes control risk. However, other explanations could potentially result in these same patterns of risk to relatives. Thus, even sophisticated statistical approaches to modeling genetic inheritance have failed to define any single biological mechanism controlling risk to orofacial clefts in all families; indeed it has been suggested that multiple genes, perhaps two to six independent genes (possibly up to 14) could determine risk to orofacial clefts22. This concept of multiple different genes involved in the etiology of orofacial clefts is increasingly borne out by the results of genomic studies, as summarized below.Still, this multifactorial threshold model adequately explains the clear gender differences in risk to CL/P and CP  and the higher risk to relatives of more severe cases Different types of genetic studies create the opportunity to map genes in the absence of a well-defined model of inheritance. There are two general types of approaches: “linkage analysis” and “association analysis”. Linkage analysis always requires multiplex families and tests for co-segregation of observed genetic markers and a hypothetical gene controlling the affected vs. unaffected phenotype within a family, while association analysis simply tests for differences in frequencies of markers in samples of affected and unaffected individuals from a population, which could reflect linkage at a population level (termed “linkage disequilibrium”). Linkage analysis is a powerful approach for mapping individual genes for traits following clear Mendelian patterns within multiplex families, but it is less effective in mapping genes for complex traits .Both linkage and association analysis can be used on one genetic marker or on many scattered over the entire genome. Importantly, both approaches can be adapted to accumulate evidence when multiple causal variants (in different genomic locations) exist in the data being analyzed (multiplex families or groups of cases and controls), and thus both can be used effectively to study orofacial clefts at the genome-wide level.23, but limits on the number and size of these families and the underlying high level of “linkage heterogeneity” (where different families yielded evidence of co-segregation to different genes) made it extremely difficult to obtain consistent results. A consortium of multiple research groups pooling genome-wide markers (~400 microsatellite markers) identified linkage to six different chromosomal regions 25. Further fine-mapping studies of these regions showed consistent evidence of linkage to chromosome 1q32 near theIRF6 gene. Interestingly, mutations inIRF6 account for most of the known cases of Van der Woude syndrome, a Mendelian autosomal dominant syndrome which includes orofacial clefts as a key phenotype23. In addition, markers in and near this gene have also consistently shown evidence of association in population-based studies. Another gene,FOXE1 on chromosome 9q21, yielded consistent evidence of linkage in multiplex families26 and recent association studies28. Thus, linkage studies based on multiplex families successfully identified some causal genes29, but clearly multiple genes control risk to orofacial clefts and these vary from family to family.Numerous groups used linkage studies to map genes in multiplex cleft familiesAssociation studies ask a much more general question than linkage analysis, and typically this involves rather simple statistical comparisons of marker allele or genotype frequencies between groups of unrelated people. While interpreting a positive result from linkage analysis represents more compelling evidence that a putative gene controls risk, these multiplex families are rare (i.e. most cleft cases do not have any affected close relative) and individual families are not representative of the full population. Interpreting statistically significant results from an association study is more ambiguous and less definitive than results from linkage analysis, and unrecognized bias in the study design can produce spurious results . As with linkage studies, one, several, or millions of markers can be compared between case and control groups, but of course as the number of tests increases some accommodation must be made for false positive results obtained by chance alone . Conversely, false negative results can become important too, i.e. where a key risk gene is missed because the markers used for tests of association are not informative and do a poor job of covering that gene in one population or another.30, but the results were not always consistent across studies. This inconsistency could merely be because of limited sample sizes of individual studies, subtle biases in study design, or could reflect genetic differences across populations. Interestingly, sequencing the candidate genes themselves (and the regions immediately surrounding the gene) has identified some rare mutations that would impair the function of the corresponding gene product and could be directly causal32. While such rare variants are potentially biologically relevant and certainly justify further study of that candidate gene, their very low frequency in the population makes them difficult to incorporate either into standard statistical tests or for use in predicting risk to individuals (which is notoriously difficult for heterogeneous disorders).Association studies using case–control designs have been used to examine numerous selected “candidate” genes for orofacial clefts (CL/P and less commonly CP) for over 25 yearsst century and building upon the Human Genome Project, genome-wide association studies (GWAS) became feasible as genotyping technology improved to allow hundreds of thousands (or millions) of single nucleotide polymorphic (SNP) markers to be typed efficiently on large samples. The first GWAS of orofacial clefts used a classic case-control design to compare markers between CL/P cases from Germany to “universal” or unphenotyped German controls33. The use of “universal” controls  is reasonable because the birth prevalence of orofacial clefts is relatively low. However, when conducting association studies, even subtle differences in the genetic background between the case and control groups can be a potential source of bias and must be considered. The key advantage of GWAS is its ability to identify simultaneously one or several regions of the genome likely to contain genes controlling risk to orofacial clefts. In the presence of multiple causal genes, there should be multiple genomic regions giving statistically significant results.Beginning early in the 2135. This design was used by Beatyet al.36 in an international collaboration supported by the Genes and Environment Association (GENEVA) consortium37 where distinct racial groups (European and Asian) were well represented.An alternative to the traditional case–control design is the family-based case–parent trio or “triad” design, which has the advantage of being robust to the confounding caused by “population stratification” or differences in marker allele frequencies between groups of cases and controls. Now the contrast is between the marker  seen in the case and the set(s) that are possible given the parental mating type, yielding a comparison between the observed case and a matched set of “pseudo-controls” drawn from the parental mating type. This study design actually tests a composite null hypothesis of strict Mendelian transmission of marker alleles (or their genotypes), and rejecting this composite null represents evidence for both linkage and association38 listed a dozen different genes/regions attaining genome-wide significance either in one GWAS or from meta-analysis across multiple GWAS, andIRF6 on 1q32–41, the gene desert region on 8q24, 17q22, and 10q25.3) appear to account for 20–25% of the estimated genetic variation in risk or “heritability” to CL/P, a much larger proportion of the estimated heritability attributable to markers identified by GWAS than seen for many other complex disorders17.To date, there have been seven independent GWAS for CL/P and two for CP. They have identified at least two dozen different genes/regions achieving genome-wide significance for CL/P, but only one for CP. Leslie and Marazita39 or true differences in the frequency of causal mutations across populations. Because this strong statistical evidence comes from SNPs in a “gene desert” with no recognized coding regions nearby, it is particularly intriguing that mouse models have suggested this region may contain a regulatory region critical for normal craniofacial development40.However, the level of statistical support is frequently limited to one or another racial/ethnic group (e.g. the support for the gene desert region in 8q24 is consistently strong in European ancestry groups but much less so among Asian ancestry groups). This may reflect limitations in “tagging” by widely used SNP marker panelsIn many ways, the progression in genetic studies of orofacial clefts has repeatedly applied standard tools of genetics to a disorder that is unexpectedly complex and heterogeneous. Clearly there is a genetic basis to the etiology of orofacial clefts; however, it must involve multiple genes (maybe dozens) and several environmental risk factors (which may interact with different genes). Our analytical tools, both statistical and technological (for genotyping and sequencing), have greatly expanded in recent years but always seem to fall short of the mark in dealing with the high level of etiologic heterogeneity for this relatively rare disorder. Technological advances now make it feasible to sequence the entire exome  and even the entire human genome. However, we still must carefully consider the most appropriate study design and how to analyze sequence data most effectively.41, but if many rare mutations can lead to the same phenotype, it is entirely possible that each family could reflect effects of a unique gene or unique combination of genes. Bureauet al.43 recently showed how whole exome sequencing (WES) data in distant affected relatives (second-degree and more distant relatives) drawn from multiplex families originally recruited for linkage studies could be used to identify rare variants that modify risk to orofacial clefts. From a long list of candidate genes, this approach identified one novel variant inCHD1 shared by three affected second cousins in a single multiplex family.CHD1 is a strong candidate gene because mutations in this gene can cause both gastric cancer and orofacial clefts, and several studies have shown evidence of association or excess of rare variants in this gene47. A comprehensive whole exome analysis revealed a separate novel variant inADAMTS9 achieving statistical significance over the entire exome in a small number of families42. While this gene had been identified as potentially causal in mouse models, this is the first evidence from human studies, and the risk allele is extremely rare. This same method of testing for rare shared variants can be applied to structural variants, such as deletions detected from WES data48.Family study designs can be very effective in identifying causal geneset al.49 demonstrated the potential of sequencing studies in a large study of case–parent trios (where 13 regions around candidate genes were sequenced). This case–parent trio design allows the detection ofde novo mutations, where a new mutation (not present in the parents) occurs in the child. This type of mutation simply cannot be identified in case–control studies, and such mutations are by their nature extremely rare. Also, if ade novo mutation lies outside any known coding or regulatory region, it is difficult to judge its true functional status. This group concludedde novo variants probably don’t represent a major etiologic component for orofacial clefts, although they may be relevant for other complex diseases.LeslieNOG andNTN1) showed some statistical evidence for distinct rare variants within the gene being important in controlling risk, while four of the 13 genes yielded no such evidence . Because obtaining statistical evidence for rare variants is always difficult, even in large studies, building a biological case for a putative causal mutation requires functional studies for confirmation49.This case–parent trio study of sequencing data yielded evidence consistent with previous GWAS studies, where multiple genomic regions showed strong statistical evidence of linkage and association for common polymorphic markers, but it also was able to analyze rare variants that could themselves be directly causal for orofacial clefts. Careful analysis of low-frequency and rare variants using available methods for collapsing or combining multiple variants within a gene to assess the overall impact of that gene on risk yielded mixed results: two of 13 candidate regions  studies are now underway and are being applied to a number of complex and heterogeneous diseases . Practical challenges in using WGS data include cleaning and processing the huge volume of data. The amount of data scales up dramatically from GWAS to WES to WGS: GWAS data represent hundreds of thousands or millions of typed markers ; WES represents 1–2% of the human genome sequenced 30–50 times and aligned to yield a consensus sequence for ~50 Mb of the genome; WGS represents similarly aligned read data on all 30 billion base pairs of the genome . The primary advantage of WGS data is that it can be used to analyze any type of sequence variant from the reference genome , small deletions (termed indels), or larger structural changes . The challenge comes in defining the particular assessment/comparison being made with variants identified by WGS data , and then interpreting any “significant” differences from expected. The study design always determines the hypothesis to be tested , but the huge number of tests must be considered, along with the potential biases inherent in the study design. Dealing with WGS data will involve tens of millions of markers or SNVs on individuals, and most of these will be quite rare. Most observed sequence variants will occur outside known coding regions, and our ability to recognize important regulatory regions remains limited.st century, and it is our challenge to use them as effectively and efficiently as possible. It is likely that there will be classes of “causal genes” and “genetic risk factors”: a few genes with “major effects” may simultaneously control truly Mendelian malformation syndromes and show association with apparent non-syndromic forms of orofacial clefts, while possibly many “genetic risk factors” have the typical modest estimated effect sizes identified through GWAS but nonetheless achieve genome-wide significance as studies are combined via meta-analysis.The more important challenge facing genetic studies of orofacial clefts is to develop strategies for dealing with multiple causal genes within any data set and applying these strategies effectively to orofacial clefts as a complex and heterogeneous birth defect. The full range of study designs  is available, but interpreting the biological function of a novel variant detected by sequencing will not be simple. Furthermore, because any given patient could represent the effects of one or more variants in several different genes, translating new genetic findings into meaningful medical or public health applications remains a daunting task. In reality, for orofacial clefts (or any other birth defect), because the first occurrence occurs well before birth, it is difficult to define opportunities to intervene. Nonetheless, the tools to detect genes controlling risk to orofacial clefts have expanded greatly in this first part of the 21CI, confidence interval; CL, cleft lip; CL/P, cleft lip with or without cleft palate; CLP, cleft lip and palate; CP, cleft palate; GWAS, genome-wide association study; GxE, gene–environment; SNP, single nucleotide polymorphism; SNV, single nucleotide variant; WES, whole exome sequencing; WGS, whole genome sequencing.


MBL2 gene encodes mannose-binding lectin, is a member of innate immune system. Earlier studies revealed that MBL2 gene variants, rs1800451, rs1800450, rs5030737, rs7096206, rs11003125 and rs7095891 are associated with impaired serum level and susceptibility to TB, but their results are inconsistent. A meta-analysis was performed by including 22 studies (7095 TB-patients and 7662 controls) and data were analyzed with respect to associations between alleles, genotypes and minor allele carriers to evaluate the potential association between MBL2 polymorphisms and TB risk. Statistically significant results were found only for the homozygous variant genotype  of rs1800451 and showed reduced risk of TB in overall population. However, other genetic models of rs1800450, rs5030737, rs7096206, rs11003125, rs7095891 and combined rs1800450, rs1800451, rs5030737 polymorphisms of MBL2 gene did not reveal any association with TB risk. Stratified analysis by ethnicity showed decreased risk of TB in African population for rs1800450 and rs1800451. Whereas, no association was observed between other MBL2 polymorphisms and TB risk in all the evaluated ethnic populations. In conclusion, MBL2 rs1800450 and rs1800451 polymorphisms play a protective role in TB infection and reinforce their critical significance as a potential genetic marker for TB resistance. Mycobacterium tuberculosis (M. tuberculosis) and still a major chronic infectious disease in most parts of the world. Despite applying the various treatment strategies, which were supposed to eliminate this infectious disease from the root, but recent data revealed that TB is once again on the upsurgeM. tuberculosis and fairly large number of population leftover with no clinical symptoms of this infectious disease. However, nearly 5–15% of the individuals will develop the active disease as confirmed by culturable bacilli from the sputum and by other clinical symptomsM. tuberculosisM. tuberculosisTuberculosis (TB) is essentially caused by MBL2 gene encodes for mannose-binding lectin (MBL) [also called mannose-binding protein or mannan-binding protein (MBP)], is mapped on 10q11.2-q21 chromosome, instrumental in innate immunity via the lectin pathway. MBL is a liver derived complement activating serum lectin, which binds specifically to the oligosaccharide structure present on various microorganisms comprising M. tuberculosis, mediates phagocytosis, and activates the mannose-binding lectin pathway of the complementMBL2 geneThe MBL2 gene, out of which six are acknowledged for their functional effects. Three SNPs are located at exon 1: first, rs5030737 is a C > T transition at codon 52 (CGT > TGT) that results in Arg52Cys substitution , second, rs1800450 is a G > A transition  at codon 54 (GGC > GAC) resulting in Gly54Asp substitution, and third, rs1800451 is a G > A transition  at codon 57 (GGA > GAA) resulting in Gly57Glu substitution. These three SNPs are responsible for the disruption of the collagenous back-bone of the MBL molecule and leads to protein dysfunctionMBL2 gene, viz. −550 C > G , −221 G > C  and +4 C > T , that can affect the transcription rate, and consequently the concentration of serum MBL. Up to 1000 fold variations in MBL concentration level have been noticed in different individuals possibly because of deviating actions caused due to the combination of structural genes and promoter polymorphismsNumerous SNPs have been identified in MBL2 gene and lead to low level of the MBL2 protein, which affects the innate immunity against the pathogen. A number of clinical (case-control) and genetic studies have been done in the past to review the impact of various MBL2 gene polymorphism on the development of TB111213141516171819202122232425262728293031MBL2 gene polymorphisms and genotype on susceptibility to TB infection development are inconsistent and inconclusive. The inconsistency in the results across many of the case-control studies could possibly be due to the ethnicity of the population under consideration, sample size, and low power of the individual studies to determine the overall effect. In order to overcome this issue, nowadays a meta-analysis based statistical tool is in use to evaluate the risk factors associated with the genetic diseases, as it employs a quantitative method of pooling of the data drawn from individual studies, where small sample sizes are insufficient to deliver precise and reliable conclusions. Therefore, the present meta-analysis was performed based on literature identification until 30 March 2016 to evaluate the effect of various MBL2 polymorphisms, rs1800451 (A > C), rs1800450 (A > B), rs5030737 (A > D), rs7096206 (Y > X), rs11003125 (H > L), rs7095891 (P > Q), and combined rs1800450, rs1800451, rs5030737 (A > O), on the risk of overall and ethnicity based effect on TB infection.Since, all the above stated polymorphisms affect MBL2 polymorphisms and TB risk were screened. After careful reading of the full-text of all the screened articles, 5 studies were disqualified . During the study selection, all the retrieved articles were reviewed cautiously, and publications either dealing with MBL2 variants to predict survival in TB patients or MBL2 polymorphisms as indicator for response against therapy were disqualified. Likewise, studies pertaining to protein expression or MBL2 mRNA levels or relevant review articles were also omitted from this meta-analysis. In this meta-analysis, only case-control or cohort design studies mentioning the frequency of all the three genotypes were included. In addition to the online database search, the references given in the screened articles were also reviewed for other potential studies. Lastly, after cautious screening and following the pre-set inclusion and exclusion criteria for the selection of the studies, a total of 22 research publications111213141516171819202122232425262728293031The preliminary search of PubMed (Medline), EMBASE and Google Scholar web-databases resulted 52 articles using the selected key words as mentioned in the methods section, and after detailed evaluation of the titles and abstracts, and after eliminating the duplicates, 27 articles dealing with evaluation of the association of MBL2 polymorphism  to evaluate the influence of each individual study on the pooled OR by eliminating each single case-control study. The results of sensitivity analysis revealed that no individual study influenced the pooled ORs significantly in all the MBL2 variants and endorsed the credibility and stability of the present meta-analysis  and TB risk, (b) should clearly designated confirmed TB patients and TB free controls, (d) have available genotype frequency in both the cases and controls to calculate odds ratio (OR) and 95% confidence interval (CI), (e) independent studies showing original data, (f) published in the English language, (g) followed statistically acceptable data collection and analysis methodology. Moreover, when the case-control study was involved in more than one research publication using the same case series, the research study that incorporated the highest number of individuals was included in this analysis. The studies were excluded based upon the following criteria: (a) duplicated or overlapping publication, (b) study design based on only TB cases, (c) without genotype distribution or allele frequency data, (d) clinical case studies without control subjects, (e) data of review, abstracts, or case reports.The identified studies in this meta-analysis had to meet all of the below given criteria to reduce heterogeneity and facilitate the apt interpretation of this study: (a) they must have evaluated unrelated case-control studies or cohort design between MBL2 polymorphisms, and reported associations. Cases of disagreement or discrepancy in any item of the abstracted data from the selected articles were resolved by conducting group discussions involving adjudicator (SAD) to accomplish a final consensus.The methodological quality evaluation and data extraction were independently performed in duplicate copies by two independent investigators (RKM & NA) using a standard process for each retrieved article, based on the pre-set eligibility criteria of the study inclusion in the present study, and sequential exclusion of the inappropriate studies. In order to confirm the accuracy of the abstracted data by strictly following the given selection criteria a standard data-collection form was used. The following data were abstracted from each included study: publication year, the name of first author, origin country, ethnicity, source of cases and controls, number of cases and controls, type of TB [pulmonary TB (PTB) or extra-pulmonary TB (EPTB)], study type, genotype frequencies, The Newcastle-Ottawa quality assessment Scale (NOS)MBL2 gene polymorphisms with the risk of developing TB. Chi-square-based Q-test was applied to test the heterogeneity assumption2 statistics was used to measure inter-study variability, and greater values indicated higher degree of heterogeneityhttps://www.meta-analysis.com/pages/comparisons.php, and finally Biostat’s  Comprehensive Meta-Analysis (CMA) Version 2 program was selected for performing all the statistical study involved in this meta-analysis.Pooled ORs and their corresponding 95% CIs were calculated to appraise the association of How to cite this article: Areeshi, M. Y. et al. A Meta-analysis of MBL2 Polymorphisms and Tuberculosis Risk. Sci. Rep.6, 35728; doi: 10.1038/srep35728 (2016).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


MTHFR gene (OMIM: 607093) plays an important role in the folate metabolism. It has been suggested that C677T (rs1801133) and A1298C (rs1801131) genetic polymorphisms in the MTHFR gene lead to the decreased activity of the methylenetetrahydrofolate reductase enzyme which may have significant effect on developing schizophrenia. We used a case-control study to establish the possible association between the C677T and the A1298C polymorphisms and susceptibility to schizophrenia in an Iranian population. The genotypes of the polymorphisms were determined using PCR-RFLP. The data were analyzed by logistic regression model. Data analysis revealed that the combination genotypes of 677CT/1298AA, 677CC/1298CC, 677TT/1298AA, 677CT/1298AC and 677CT/1298CC increase the risk of schizophrenia. In order to evaluate the effect of combined genotypes of the three mentioned polymorphic loci, the frequencies of the compound genotypes were compared between control and patient groups (Table 4). Base on the results, the existence of >4 risk factors showed about 32-fold increased risk for schizophrenia .Schizophrenia is a complex disorder with polygenic inheritance. The  Schizophrenia is a complex psychotic disorder with a multiple gene inheritance, which affects approximately 1% of the general population worldwide .Folate iMTHFR (OMIM: 607093) gene is located at 1p36.3. Positive linkage of 1p36 with schizophrenia has been reported [MTHFR gene, C677T (rs1801133) in exon 4 and an A1298C (rs1801131) in exon 7 are functional and result in diminished enzyme activity [MTHFR in schizophrenia has been confirmed by the observation of clinical improvement in this psychiatric illness with folate [MTHFR 677TT genotype with low plasma folate concentrations compared to subjects with the normal genotype [MTHFR polymorphisms and schizophrenia, but the results are not consistent [The FR OMIM: 07093 genC677T rs101133 in h folate -16. Highgenotype . Several C677T rs01133 in nsistent .GRIN1; OMIM:138249) plays a fundamental role in many brain functions and its involvement in the pathogenesis of schizophrenia has been widely investigated [GRIN1 and developing of schizophrenia [The pathogenic mechanisms by which the hyperhomocysteinemia causes diseases are not fully understood, but there is an evidence that homocysteine disrupts the neurodevelopment by acting as an antagonist of the N-methyl-D-aspartate (NMDA) receptor glycine co-agonist site . Furtherstigated . Recentlophrenia .MTHFR polymorphisms and susceptibility to schizophrenia was investigated. According to possible effect of both MTHFR polymorphisms and the G1001C variant of GRIN1 on NMDA receptor activity, we also investigated combination of these polymorphisms as a potential risk factor for developing schizophrenia.In the present study, association between Subjects: Two hundred  unrelated patients with a mean age of 43.3 (SD = 11.3) were recruited from hospitals in south and southwest Iran. The control group consisted of 200 healthy blood donors  with a mean age of 39.4 (SD = 11.1), which were matched for gender and ethnicity to the patient. Informed consent was obtained from all participants.Genotyping assay: Genomic DNA was extracted from leucocytes using standard salting out method. The C677T and A1298C polymorphisms in MTHFR gene were determined by PCR-RFLP using HinfI and MboII restriction enzymes, respectively. To confirm the results, a random selection of 20% of all samples was retested. There were no discrepancies in replicating test. The evaluation of GRIN1 polymorphism has been explained in details and reported before [d before .Statistical analysis: Comparison of genotypes between gender groups was done by chi square-test. The association between genotypes and development of schizophrenia was examined using odds ratios (ORs) and 95% confidence intervals (CIs). Statistical significance was considered at P< 0.05.The study of socio-demographic features of this case-control study identified that subjects in the case group were significantly lower than those in the control group in terms of marital status and educational level .Control and patient groups were initially classified according to their gender. Considering that there was no statistical difference between their genotypic frequencies (P>0.05), the samples were pooled. Detailed results of genotyping assay for C677T polymorphism are shown in In our samples, neither heterozygosity  nor homozygosity  of the T allele was associated with the increasing risk of schizophrenia . There wWe next studied the joint effect of the two polymorphisms . The com>4 risk factors showed about 32-fold increased risk for schizophrenia .In order to evaluate the effect of combined genotypes of the three mentioned polymorphic loci, the frequencies of the compound genotypes were compared between control and patient groups . Base onIn the present investigation, we did not find any significant association between the C677T polymorphism and schizophrenia. Our result is in accordance with some other investigations , 23, 34.MTHFR gene, the A1298C, reduces enzyme activity for about 30-40% [The second polymorphism in the t 30-40% , 35 and t 30-40% , 26. In t 30-40% , 23. Thet 30-40% , 36, 37.MTHFR on developing schizophrenia [Previous studies postulated a combined association between the A1298C polymorphism and the C677T variant with MTHFR activity and homocysteine levels . Althougophrenia .de novo biosynthesis of the methyl group using folate-dependent one-carbon pathway [Low contents of the SAM in human tissues could result from deﬁciency in the  pathway , 40. The pathway . The C67 pathway , 12. On  pathway  which ep pathway , 43. DNA pathway . Reduced pathway . In the  pathway .Hyperhomocysteinemia can result from decreased MTHFR enzyme activity, owing to genetic polymorphisms . ReglandTo our knowledge, the present study is the first to evaluate this combination effect. Interestingly, we found a significant association between the coexistence of these polymorphisms and increase the risk of schizophrenia.


Single nucleotide polymorphisms (SNPs) in three microRNAs (miRNAs), rs2910164 in miR-146a, rs11614913 in miR-196a2, and rs3746444 in miR-499, have been associated with breast cancer (BC) susceptibility, but the evidence is conflicting. To obtain a more robust assessment of the association between these miRNA variants and BC risk, we carried out a meta-analysis through systematic literature retrieval from the PubMed and Embase databases. A total of 9 case-control studies on rs2910164, 12 on rs11614913, and 7 on rs3746444 were included. Pooled odds ratios and 95% confidence intervals were used to evaluate associations with BC risk. Overall analysis showed that rs2910164 was not associated with BC susceptibility in any genetic model, whereas rs11614913 was associated with a decreased risk in both the allelic contrast and recessive models, and rs3746444 imparted an increased risk in all genetic models. Stratified analyses showed that rs11614913 may decrease the risk of BC in the heterozygote model in Asians, and in all genetic models, except the heterozygote model, when the sample size is ≥ 500. Subgroup analysis indicated that rs3746444 was associated with increased risk of BC in Asians, but not Caucasians, at all sample sizes. This meta-analysis suggests that rs11614913 in miR-196a2 may decrease the risk of BC, while rs3746444 in miR-499 may increase it, especially in Asians when the sample size is large. We propose that rs11614913(C > T) and rs3746444 (A > G) may be useful biomarkers predictive of BC risk. One of the most surprising advances in understanding the mechanisms of gene regulation in health and disease has been the discovery of microRNA (miRNA) .MiRNAs are short , evolutionarily conserved, noncoding RNA molecules that exert post-transcriptional regulation via binding to complementary sequences in the 3′-untranslated region (3′ UTR) or 5′-untranslated region (5′UTR) of target messenger RNAs (mRNAs) . Thus, mBreast cancer (BC) is the most common type of cancer affecting women worldwide. Although its etiology is multifactorial, its development and outcome are especially influenced by genetic factors. In this regard, several studies pointed out that alterations in miRNAs may contribute to the pathogenesis of BC , 7. SincSingle nucleotide polymorphisms (SNPs) represent the most common form of genetic variation. When present in miRNA genes, SNPs may influence miRNAs’ properties by altering their expression, maturation, and/or function , and mayAmong several common miRNA SNPs purportedly related with BC risk, the association of rs2910164 in miR-146a, rs11614913 in miR-196a2, and rs3746444 in miR-499 and BC risk remains inconclusive. For instance, Bansal et al. found that the heterozygous variant of rs2910164 in miR-146a is associated with reduced risk of BC , while aBased on the search strategy, a total of 217 articles were chosen from PubMed and EMBASE databases. After screening the title and abstract, 192 articles uncorrelated with BC risk and miR-146a/-196a2/-499 SNPs were excluded. 25 articles were then evaluated in detail, and 8 articles were further excluded, among which 3 were meeting reports –18, 2 haQ-test and 2I statistic under all genetic models except the heterozygote. Therefore, except for the latter, the random-effects model was used for all models. No significant associations were identified for any genetic model   Figure , Table 2Next, subgroup analysis was carried out according to race. No significant association was found in any genetic model for Asians and Caucasians. Subgroup analysis based on the source of controls revealed no significant association between any genetic model and either population-based or hospital-based groups; also, no associations were detected for sample sizes < 500 and ≥ 500 Table .The association between miR-196a2 rs11614913 polymorphism and the risk of BC was tested using the random-effects model, due to the presence of significant heterogeneity, for the allelic contrast model and the homozygote, dominant, and recessive models, while the fixed-effects model was used for the heterozygote model. A significantly decreased risk of BC was observed under the allelic contrast model and the recessive model . In Caucasians, in contrast, no association was detected between miR-196a2 rs11614913 and BC risk for any genotype model. Similarly, no relationship was found for any genotype model in the mixed-race subgroup. Results of subgroup analysis based on the source of controls showed no significant association between any genetic model and either population-based or hospital-based controls. We also found no significant association for sample size < 500 under any genetic model, although a sample size ≥ 500 was associated with decreased BC risk in all models except the heterozygote . In contrast, no association was found for Caucasians under any genetic model. Subgroup analysis by source of controls indicated an association with hospital-based controls under all genetic models . Also, a significant association was identified between all genetic models and sample size ≥ 500 , and between the dominant and heterozygote models and sample size < 500 , heterozygote , dominant  and recessive  models.We utilized Begg's funnel plot and Egger's test to evaluate publication bias. No evidence of publication bias was found for the association between miR-146a rs2910164, miR-196a2 rs11614913, and miR-499 rs3746444 polymorphisms and BC susceptibility using Begg's funnel for the allelic contrast model  polymorphism in miR-146a may not be associated with susceptibility to BC; the rs11614913 (C > T) polymorphism in miR-196a2 is significantly associated with decreased BC risk; and the rs3746444 (A > G) polymorphism in miR-499 is associated with increased BC risk, especially in Asians. Thus, rs11614913(C > T) and rs3746444 (A > G) appear to be both promising biomarkers to forecast BC risk and potential therapeutic targets. However, owing to the limitations mentioned above, these results should be treated with caution. To further verify and confirm these findings, well-designed, large scale case–control studies will be required.This meta-analysis was performed in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement . To idenP > 0.05). Exclusion criteria included: (1) duplications of previous publications; (2) comments, meeting reports, reviews or editorials; (3) family-based studies of pedigrees; (4) studies with no detailed genotype data. When there were multiple publications from the same population, only the largest study was included. Study selection was done by two investigators independently, by screening the title, abstract and full-text. Any dispute was settled by discussion.All the included studies were selected following the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) guidelines . EligiblP value for HWE (Table Data from eligible studies were extracted in duplicate by two investigators independently (Mu and Guo). Extracted data included author, year, country, race, genotyping method, source of control, genetic models of cases and controls, and WE Table . These tP value of HWE in the control group using an online tool (http://ihg.gsf.de/cgi-bin/hw/hwa1.pl). The departure from HWE of SNP frequencies in the control group was assessed by 2X test, and a P value < 0.05 was regarded as significant. Odds ratios (ORs) and 95% confidence intervals (CIs) were obtained to evaluate the strength of the association between miR-146a/-196a2/-499 SNPs and susceptibility to BC. Pooled ORs were determined for the allelic contrast model , homozygote model , heterozygote model , dominant model , and recessive model . The statistical significance of the pooled OR was evaluated by Z test and a P value of < 0.05 was regarded as significant. Inter-study heterogeneity was tested using a 2X-based Q-test (with significance level P < 0.1) and 2I statistic  [P < 0.05; conversely, the fixed model was selected when P > 0.05. Subgroup analysis was performed by race, source of control, and sample size. Sensitivity analysis was performed to evaluate the effect of each study on the combined ORs by omitting individual studies one at a time. Publication bias was checked by Begg's funnel plots [P < 0.05 for the Egger's test denoted a noteworthy publication bias. The trim-and-fill computation was used to estimate the effect of publication bias on the interpretation of the results [We calculated the geneity) . Accordiel plots  and Eggeel plots . An asym results . Statist


MTHFR) c.677C>T and c.1298A>C variants were known to be associated with prostate cancer (PCa) risk with conflicting results, because of MTHFR and nutrient status interaction in the prostate development. In this large-scale, hospital-based, case-control study of 1817 PCa cases and 2026 cancer-free controls, we aimed to clarify the association between these two MTHFR variants and PCa risk in Shanghai and to explore the underlying molecular mechanisms. We found that both the heterozygous CT  and the homozygous TT genotypes  of c.677C>T were associated with a significantly decreased risk of PCa compared with homozygous wild-type CC genotype, respectively, using multivariate logistic regression. Furthermore, we confirmed that MTHFR c.677T allele was related to an increased serum homocysteine level in the Han Chinese population in Shanghai. In the cultured PCa cell lines, we observed that MTHFR c.677T could elevate the cellular homocysteine level and cause DNA damage, thus increasing cell apoptosis and finally inhibiting cell proliferation. In conclusion, MTHFR c.677T was a protective factor of PCa risk in ethnic Han Chinese males by inducing DNA damage and cell apoptosis.Methylenetetrahydrofolate reductase ( Prostate cancer (PCa) is the second most frequently diagnosed cancer and the fifth leading cause of cancer death worldwide among men, and remains the first frequently diagnosed cancer in more developed countriesPCa is a complex disease; and many etiological factors, including genetic profile, nutrition, environmental exposures, etc., are thought to play important roles in cancer development78MTHFR) gene encodes an enzyme that catalyzes the irreversible conversion of 5′,10′-methylenetetrahydrofolate to 5-methyltetrahydro folate, which serves as a methyl donor in the remethylation of homocysteine to methionine. Until now, the MTHFR gene has been identified to possess 98 different variants with benign, pathogenic or unknown significance which are listed on ClinVar database (http://www.ncbi.nlm.nih.gov/clinvar/). Among these variants, rs1801133 (c. 677C>T) and rs1801131 (c.1298A>C) are two most extensively reported polymorphisms with a global minor allele frequency (GMAP) 0.245 and 0.249, respectively19212223Folates are cofactors and cosubstrates for biological methylation and nucleic acid synthesis and also act as regulatory molecules. The folate-mediated, one-carbon metabolic pathway is fundamental to DNA synthesis, repair and methylation17et al. reported a significant reduction of PCa risk in yellow people who carried c.677T variant in Taiwan26et al. performed study in Taiwan while Cai et al. performed study in the Liaoning Province, both far different regions from Shanghai. There was a limited sample size in the previous study regarding MTFHR variants in Han Chinese PCa susceptibility; and more importantly, this association has not been surveyed in Shanghai city, which had the highest PCa incidence in China. The current study aimed to investigate the association between these two common variants of MTHFR and PCa risk in a large-scale, hospital-based, case-control cohort in Shanghai city and to explore the underlying molecular mechanisms.Wu The demographic characteristics of the 1817 PCa patients and 2026 normal controls were presented in MTHFR gene SNPs among controls met the criteria of Hardy-Weinberg Equilibrium . We found a significantly different genotype frequency distribution of rs1801133 (C677T) between the PCa group and the cancer-free group (P < 0.001), while no different genotype frequency distribution of rs1801131 (A1298C) was observed between cases and controls (P = 0.690). Both the heterozygous CT genotype  and the homozygous TT genotype  of rs1801133 were associated with a significantly decreased risk of PCa, compared with homozygous wild-type CC genotype, respectively. In addition, the relationship between the two MTHFR genotypes and the risk of PCa was evaluated using dominant models, recessive models and additive models, respectively, with all three models achieving statistical significance; which indicated that people who carried T allele of rs1801133 had a lower risk of developing PCa. Besides, no clear association between genotype frequency of rs1801131 and PCa risk was observed. Furthermore, we used multivariate logistic regression analyses to reanalyze the above-mentioned genotypes after adjusting for age at diagnosis, BMI, hypertension, diabetes mellitus, and cardiovascular disease. We found that all of the genotypes maintained the same relationship with PCa risk as declared above in a univariate logistic regression model. Finally, we found a 32% decreased risk of PCa for people who carried homozygous TT genotype of rs1801133  compared with those who carried homozygous wild-type CC genotype.The genotype frequencies and their associations with the risk of PCa were summarized in 2-based Q test . Besides, a decreased risk of PCa was observed among subgroups age >68 years, a BMI of <25 kg/m2, with hypertension, without diabetes mellitus and without cardiovascular disease for patients carrying MTHFR rs1801133 TT genotype; and an increased risk of PCa was observed among subgroups with cardiovascular disease with lymph node involvement and those carrying the MTHFR rs1801131 CC genotype, further homogeneity tests didn’t support the differences in PCa risk found between these strata.We performed a stratified analysis to evaluate the association between SNPs mentioned above and PCa risk in different subgroups using a recessive model. Analytic results were presented in MTHFR may alter enzymatic function and change the serum homocysteine level. To test this hypothesis, we randomly selected 306 participants from a control group and detected the serum homocysteine levels. Among the 306 participants, 89 (29.1%) carried the wild-type CC genotype of rs1801133, 147 (48.0%) carried the heterozygous CT genotype and 70 (22.9%) carried the homozygous TT genotype. We found that people with the heterozygous CT genotype or homozygous TT genotype had a significantly higher serum homocysteine level compared with wild-type CC carriers . Overexpression of c.1298A and c.1298C MTHFR in MTHFR-knockdown cell lines both could repair DNA damage (data not shown).We performed Comet Assay to assess DNA damage. In normal cells, the fluorescence is confined mostly to the nucleus because undamaged DNA cannot migrate. In cells with DNA damage, DNA is denatured by the alkali solution used for single-strand break detection or the neutral solution used for double-strand break detection. The negatively charged DNA fragments are then released from the nucleus and migrate toward the anode. The results of control cells and MTHFR-knockdown, shMTHFR + wild-type MTHFR and shMTHFR + c.677T/c.1298C MTHFR) in LNCaP and PC3 cell lines, respectively. We found that the c.677T variant of MTHFR significantly increased cell apoptosis compared with wild-type genotype in both cell lines  will cause about a 70% and 40% reduction of enzyme function compared with wild-type genotype, respectivelyMTHFR gene was a protective factor of PCa risk. However, some research groups found that the 677TT genotype of the MTHFR gene might allow the expression of tumor suppressor genes and reduce the risk of cancer development while it partly caused enzymatic dysfunctionMTHFR inhibition was proven to arrest the growth of cancer cells in in vitro studies because of limited methionine supplyMTHFR variants in PCa carcinogenesis, development, and progression.It has long been recognized that deficiency of folate may increase the incidence of cancer and many other diseases. Deficiency of folate contributes to a higher homocysteine level, thus leading to increased cytosolic calcium and reactive oxygen species (ROS)MTHFR polymorphism on PCa risk, was another condition not available from participants. Finally, potential mechanisms of the long-distance regulatory effect of the two studied MTHFR polymorphisms on the expression of MTHFR was not explored in-depth, which warrants further investigation in our future study. Notwithstanding these limitations, our research underscores that ethnic Han Chinese males carrying MTHFR c.667T allele had a significantly lower risk for developing PCa compared with those with wild-type MTHFR genotype of rs1801133. Further functional analyses revealed that the MTHFR c.677T contributes to elevating homocysteine level, increasing DNA damage and cell apoptosis, and inhibiting proliferation of PCa cells, thereby decreasing PCa risk.We acknowledged that there were some limitations in our study design and performance. Firstly, there may be some biases from samples collection in this hospital-based, case-controlled study. Although these biases could be minimized by taking into consideration confounding factors and age-matching between cases and controls. Secondly, in the stratified analysis, the case number in subgroups was relatively small and it would decrease the statistical power to examine the genetic variants with regard to PCa risk. Thirdly, folate intake, which is an important factor affecting the function of MTHFR c.677T variant acts as a protective factor of PCa risk by elevating homocysteine level, increasing DNA damage and cell apoptosis, and inhibiting proliferation of PCa cells. Further research focusing on gene-gene or gene-environment interactions should be conducted to explore more legible mechanisms of the role MTHFR plays in PCa risk.In conclusion, our findings indicated that This hospital-based, case-controlled study recruited 1817 newly diagnosed PCa patients, and 2026 matched cancer-null controls from genetically unrelated ethnic Han Chinese men treated at Fudan University Shanghai Cancer Center from January 2008 to June 2015. Two well-trained pathologists, specializing in genitourinary cancer, assessed the histological type of the PCa patients independently. All of the cases were defined as primary prostate adenocarcinoma according to the WHO criteria for PCa. We excluded patients with malignancies other than primary PCa, patients with a family history of PCa, and patients who received chemotherapy or radiotherapy before enrollment. Tumor stage was assessed and classified according to the TNM classification system provided by the American Joint Committee on Cancer (AJCC) in 2010Frequency-match was performed in the 2026 cancer-null Chinese men by age at diagnosis and geographic area according to the PCa patients’ characteristics. Since serum PSA screening and digital rectal examination were not routinely performed in China, we advised people with urinary tract symptoms to have a serum PSA measurement and a digital rectal examination. Participants with a serum PSA level >4 ng/ml, with or without an abnormal digital rectal examination, were also excluded from the control database.All of the study designs and testing procedures were performed according to the ethical standards of the Helsinki Declaration II and approved by the Scientific and Ethical Committee of Fudan University Shanghai Cancer Center. Written informed consents were obtained from all participants before any study-specific investigation was performed.MTHFR gene, rs1801133  and rs1801131 , were all located in exon region and were genotyped by SNaPshot analysis (ABI). Among all of the testing results, we used direct dye terminator sequencing of PCR to validate 5% of them according to the manufacturer’s instructions of ABI Prism BigDye system . The samples for sequencing and genotyping were run on an ABI 3730 automated sequencer and analyzed by SeqMan and Peakscan, respectively.We extracted genomic DNA from peripheral leukocytes using the Qiagen Blood DNA Mini Kit  following the standard protocols. Our selected SNPs of 2 at 37 °C and the media was replaced every other day.Human LNCaP and PC3 cell lines  were cultured in RPMI 1640 media  supplemented with 10% fetal bovine serum (Invitrogen). The cells were maintained in 5% COMTHFR cDNA was amplified by PCR, and CDS was cloned into the pcDNA3.1 vector . The corresponding C677T and A1298C mutated plasmid was generated by site-directed mutagenesis with the MutanBEST Kit  to ensure a uniform backbone sequence. We verified all recombinant clones by DNA sequence. The expression plasmids were transfected into cells using Lipofectamine 2000 (Invitrogen) according to the manufacturer’s instructions.Full-length MTHFR  and anti-β-ACTIN .Standard procedures were followed for western blot analysis. Antibodies used for western blotting included anti-TM siRNAs platform. The shRNA against 3′ UTR of human MTHFR gene was generated using the following target sequence: 5′-CAGTGGCAGTGAGAGCTCCAAAGAT-3′.Lentiviral vector (pLKO.1) expressing shRNA clones were generated by the Stealth RNAi7 cells were seeded in a 10 cm dish with 7 ml viral medium containing 8 μg/ml Polybrene for lentiviral infection. Cells were selected for infection by the addition of 2 μg/ml puromycin after 24 hours. Uninfected cells demonstrated 100% cell death after cultured with puromycin within 3 days.1 × 10We randomly selected 306 controls to test the circulating homocysteine level. EDTA-plasma samples were collected, centrifuged immediately, and stored at −80 °C for analyzing. Plasma and cellular homocysteine levels were detected using a tHcy Detection kit  by AU5800 biochemistry analyzer  according to the manufacturer’s protocol.The Comet Assay Kit  was used to detect single- and double-stranded DNA breaks in cultured cells. Slides were viewed (excitation 425–500 nm) with a Leica DMI 4000B epifluorescence microscope. These slides were used for each condition. In normal cells, the fluorescence is confined mostly to the nucleus because undamaged DNA cannot migrate. In cells with DNA damage, DNA is denatured by the alkali solution used for single-strand break detection, or the neutral solution used for double-strand break detection; and the negatively charged DNA fragments are released from the nucleus and migrate toward the anode.Cells were harvested and stained with Annexin V-FITC and propidium iodide  for 15 minutes at room temperature and protected from light. Finally, the cell suspension was filtered by a nylon sieve and analyzed by an Accuri C6 flow cytometer (BD Biosciences).2 incubator at 37 °C for 2 hours. Then, samples taken from each well were measured at 450 nm, and the percentage of surviving cells in each treatment group was plotted relative to the untreated one.Cell proliferation after transfection was measured using the Cell Counting Kit-8 . In brief, LNCaP and PC3 cells were seeded in 96-well plates and allowed to adhere. A 10 μl CCK-8 solution was added to each well and incubated in a humidified CO2) was calculated as the weight in kilograms divided by the height in square meters. We used the WHO cut point as an index for defining overweight status2) in the Asian population. Continuous variables were presented as mean ± standard deviation (SD) and categorical variables were reported in the form of a number (proportion).The body mass index  for evaluating genotype distribution in control subjects, and P value < 0.05 was considered deviated from equilibrium. We used univariate and multivariate unconditional logistic regression models to calculate crude and adjusted odds ratios (ORs) and 95% confidence intervals (CIs), respectively, to assess associations between the genotypes and PCa risk. In a multivariate model, covariants included age at diagnosis, BMI, hypertension, diabetes mellitus, and cardiovascular disease. Furthermore, we performed a stratified analysis to explore the relationship between the genotypes and risk of PCa among subgroups of age at diagnosis (≤68 vs. >68), BMI (<25 vs. ≥25), hypertension, diabetes mellitus, cardiovascular disease, Gleason score, extracapsular extension, seminal vesicle invasion, positive surgical margin and lymph node involvement. We used Chi-square-based Q test to evaluate the homogeneity of associations between subgroups. In all statistical analyses, a two-sided P < 0.05 was viewed as significant. We performed all statistical analyses using SPSS software version 16.0 .The goodness-of fit χHow to cite this article: Wu, J.-L. et al. MTHFR c.677C>T Inhibits Cell Proliferation and Decreases Prostate Cancer Susceptibility in the Han Chinese Population in Shanghai. Sci. Rep. 6, 36290; doi: 10.1038/srep36290 (2016).Publisher’s note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.