TNF-α -308G/A polymorphism and idiopathic recurrent miscarriage (IRM). Though three meta-analyses have been conducted on this topic, the conclusions were contradictory, and the results may be unreliable as certain crucial conditions were neglected.Conflicting results were reported on the association between the A complete search was conducted in PubMed, Cochrane Library, and Embase, other sources like Google Scholar, ClinicalTrial.gov and reference lists of relevant articles were also retrieved. All candidate articles were accessed and screened using specific inclusion and exclusion criteria. Statistical analyses were performed on data extracted from eligible studies using the STATA 12.0 software and the TSA 0.9 beta software.TNF-α -308G/A polymorphism and IRM risk. However, significant association was shown in Asian population (four studies from three publications) in the dominant model (AA + GA vs. GG), the allelic model (A vs. G), and the heterozygote model (GA vs. GG).Eventually, 12 case-control studies from 11 publications  were included in this meta-analysis, and no evidence of any significant association was found in the overall analyses between the TNF-α -308G/A polymorphism is not associated with IRM risk. Though significant association was found in Asian population, the result needs further confirmation from more studies. Some reTNF-α -308G/A (rs1800629) is most studied [TNF-α -308G/A polymorphism and IRM through a comprehensive and robust meta-analysis.Among all the genetic variants in this region,  studied –50. Howe studied , another studied , and the studied , 40, and studied –47. Mean studied  in the p studied , 41, 42 The present meta-analysis was conducted in accordance with the PRISMA  guidance . No reviRelevant articles were identified by a comprehensive search of the following electronic databases through July 2016: PubMed, Cochrane Library, Embase. The search terms included the synonyms of miscarriage, tumor necrosis factor and polymorphism . The SNPFor all the relevant literature, the following inclusion criteria were adopted: 1) case-control designed studies or retrospective cohort studies with clear inclusion criteria; 2) data on allele and genotype frequencies provided; and 3) information on DNA genotyping method and characteristics of cases and controls included. Studies without genotype data or with duplicate data were excluded. Letters, case reports, editorials, review articles, conference abstracts, and animal studies were also excluded. Eligible studies were selected by the same two authors independently by screening the title, abstract, and full article based on the above criteria. Disputes were solved by consultation.p <0.05 were considered as significantly deviating from the HWE and would be excluded from this meta-analysis. If a study had subgroups, each subgroup would be listed as a separate study. Two authors completed the whole process independently. If differences existed, data would be rechecked independently by the two authors. Further discrepancies would be referred to a third author. To obtain necessary missing data, authors were contacted via e-mail.From all eligible studies, the following data were extracted: last name of the first author, publication date, country, ethnicity, mean age and source of the cases and controls, total sample size, genotype frequencies, and genotyping method. For each study, the HWE of the control group was computed from the genotype frequencies extracted above, and studies with The quality of each study was assessed using the assessment scale adapted from Peng et al.  for the p value < 0.05 was considered significant. Heterogeneity among studies was computed by the Q statistic and the I2 statistic. For each study, either the fixed-effects model or the random-effects model was used, based on the PQ value. If the PQ value was >0.1, the former was adopted; otherwise, the latter was chosen. To investigate the influence of primary characteristics and explore the source of heterogeneity, we conducted a series of subgroup analyses. In addition, we conducted a sensitive analysis to test the stability of the overall results by sequentially taking out one study each time, a cumulative meta-analysis to portray the shift of the association over time by adding studies one by one based on publication date, and a trial sequential analysis (TSA) to minimize the risk of type I errors. Furthermore, we performed Galbraith plot to facilitate the examination of heterogeneous studies. For the evaluation of publication bias, Egger’s regression test were performed. Funnel plots and Egger’s publication-bias plots were also generated in the process. All analyses were conducted using STATA software version 12.0 and TSA software version 0.9 beta. Two-tailed p values <0.05 were considered as statistically significant.Based on the genotype frequencies of cases and controls in each study, we conducted a series of overall meta-analyses using the following five genetic models: the homozygote model (AA vs. GG), the heterozygote model (GA vs. GG), the recessive model (AA vs. GA + GG), the dominant model (AA + GA vs. GG), and the allelic model (A vs. G). Then, using the odd ratio (OR) and confidence interval (CI) produced, we evaluated the association between the above genetic models and IRM risk. The overall significance of the association was calculated by a paired z-test, and a The initial search generated 162 relevant records, of which 32 were duplicates. After reviewing the abstracts of the remaining 130 records, 104 records were ruled out as irrelevant articles, reviews, letters or case-reports. The full texts of the left 26 potential publications were obtained and reviewed. Among them, three publications without sufficient data –47, one TNF-α -308G/A and IRM risk , whereas 2 genetic models  with 11 studies available for analysis demonstrated little heterogeneity  and 95% CIs were not significantly affected, no matter which study was taken out. .TNF-α -308G/A and IRM ever appeared over time  and the allelic model (I2 29.1), indicating the four outliners may be the cause of the heterogeneity . One possible explanation is that either the cases or the controls in this study are not representative. Another indication is that the heterogeneity of the total-sample-size >150 group was significantly higher than that of its counterpart group.What is more, we conducted several stratified analyses to trace the possible sources of the heterogeneity and found several points noteworthy. First, as for the definition of IRM, no significant difference was shown between the ≥2 miscarriages group and the ≥3 miscarriages group , indicatogeneity . Fourth,More importantly, during this meta-analysis, we summarized several meaningful points that may be helpful to further studies. First, the inclusion and exclusion criteria must be specific and comprehensive and should be strictly carried out. Otherwise, confounding factors will inevitably be mixed in. Second, more studies on Asian populations are needed. Third, more attention should be paid to the selection of the control group, which is always neglected and carried out without strictly following criteria. Four studies –42 deviaThere are some limitations in the present meta-analysis, which should be noted. First, misclassification bias and selection bias may be incurred due to unified diagnostic criteria of IRM and various sources of controls. Second, we failed to evaluate the gene-gene and gene-environment associations due to lack of the original data. Third, heterogeneity in several subgroups remains high in the subgroup analyses. Last but not least, the limited number of studies in Asian population may restrict the statistical power of the association.TNF-α -308G/A promoter polymorphism and IRM was found in the present meta-analysis. The correlation found in Asian population needs confirmation from more studies.In summary, no association between the TNF-α -308G/A polymorphism and IRM risk, and the association found in the previous meta-analyses may result from the inclusion of studies inconsistent with HWE. Significant association demonstrated in Asian subjects in the subgroup analyses, should be interpreted with caution due to limited studies. Further rigorously-designed large-scale studies on Asian population are needed to confirm this conclusion.The present meta-analysis demonstrated no association between S1 Fig(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S3 Fig(TIF)Click here for additional data file.S4 Fig(TIF)Click here for additional data file.S1 File(PDF)Click here for additional data file.S2 File(PDF)Click here for additional data file.S3 File(PDF)Click here for additional data file.S4 File(PDF)Click here for additional data file.S5 File(PDF)Click here for additional data file.S1 Table(DOC)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(DOCX)Click here for additional data file.


With meta-analysis we tented to reveal the potential relationship between daily fluid consumption and bladder cancer risk, and to find out a recommendation on daily fluid intake. Databases of the Web of Science, PubMed and EMBASE were searched then 21 case-control and 5 cohort studies were included. Stratified analyses on gender, region, time of subjects recruiting and fluid quantity were performed as well as dose-response meta-analysis. Comparing the highest exposure category with the lowest in each study, no association appeared when all data pooled together (p=0.50), but a significant OR of 1.46  was found in male subgroup. For different regions, the summarized OR was 1.44 (1.10-1.89) in American case-control studies, 1.87 (1.20-2.90) in European male subgroup and 0.24 (0.10-0.60) in Asia. There was a significant relationship that each increment 1000ml daily consumption would increase the risk by 28.6% in European male (p=0.007). Similarly every additional 1000ml consumption may increase the OR by 14.9% in American people but the association wasn't that strong (p=0.057). Stratified analyses showed fluid consumption over 3000ml/day in American residents and 2000ml/day in European male resulted in OR>1 with statistical significance. In conclusion, a relationship between higher fluid intake and higher bladder cancer risk was observed in European male and American residents and a limitation to <2000ml and <3000ml per day are recommended respectively. Bladder cancer is known as one of the most common malignant tumors and the risk of bladder cancer is believed to be tightly associated with life style, fluid consumption included. However, to date there still exist inconsistent conclusion on the effect of total fluid consumption on the risk. On the one hand, the urogenous contact hypothesis states that high total fluid intake can provided protective effect against bladder cancer. It believes increased fluid intake would lead to increased urine volume and urination frequency, resulting in reducing concentration of potential carcinogens in urine as well as their bladder contact time –3. ThereTo solve this controversy, a pooled analysis and a meta-analysis were conducted in 2006 and 2014 respectively . The forTo explored more details on this problem, we conducted a meta-analysis which added missing or updated data to the previous meta-analyses. Also we conducted more specific subgroup analyses, aiming to again assess the relationship between total fluid intake and bladder cancer risk and try to conclude recommended fluid intake volumes for different population.Using the predefined search strategies, 26 article were ultimately included in the current study , 8–32, i2=84%).Risk estimates for high versus low level of daily fluid consumption are shown in Figure The outcomes of the relationship between high-level daily fluid consumption and risk of cancer in 2 gender subgroups are shown in Figure Subgroups could be divided into America, Europe and Asia according to the included studies. The summary OR of American case-control studies was 1.44  with significant association. However when cohort study pooled in, the statistic significance disappeared . For European countries, there seemed to be a higher risk in people with high-level daily fluid consumption but the association was relatively weak . And in Asia, the summary OR was 0.78  in China and 0.09  in Pakistan. For all studies conducted in Asia, the pooled OR was 0.24 .Risk estimates for the highest versus the lowest level of total fluid consumption by recruit time subgroup are shown in The outcomes above found that significant results were only observed among case control studies, and the pooled results were different among countries and genders. So further analyses were done for possible higher risk regions.As the outcomes showed, the summary OR in European female was 0.98  and 1.87  in male. Significant association only existed in male.Since positive pooled outcome showed up in case-control studies among European males, we did a stratified analysis by daily fluid intake quantity level and dose-response meta-analysis for this population (5 studies with 1973 cases and 2386 controls). According to the included studies, daily quantity (except the reference quantity) could be divide in to the following levels: the highest quantity (>3000ml), the 2nd highest quantity (2000-3000ml) and the 3rd highest quantity (<2000ml). The summary OR of these three subgroups were 2.49 , 1.67  and 1.14  respectively. Significant association existed when daily quantity reached over 2000ml Figure  & 4e.The dose-response meta-analysis included 4 researches (1633 cases and 2046 controls) and showed that association between cancer risk and fluid consumption matched a liner relation (p=0.007). Each increment 1000ml daily consumption would increase the risk by 28.6% . No evidences of non-linear relationship existed. (p=0.438) Figure .The summary OR was 1.14  in American female and 0.83  in America male. No statistic significance existed in both gender.Since American case-control studies showed higher risk in high fluid intake population, we did a stratified analysis by daily fluid intake quantity level using the data of both genders from case control studies (6 studies with 4636 cases and 6274 controls). Since upper and lower boundaries in each category varied greatly in included studies, daily quantity was divided into the following levels according to the midpoint value of each category: the highest quantity (>3000ml), the 2nd highest quantity (2500-3000ml) and the 3rd highest quantity (<2500ml). The summary OR of these three subgroups were 1.60 , 1.24  and 1.07  respectively Figure . SignifiDose-response meta-analysis of these 6 studies showed no evidence of nonlinear relationship (p=0.772). Linear analysis suggested that every additional 1000ml consumption would increase the OR by 14.9%  but without statistic significance (p=0.057) Figure .There was no asymmetry in funnel plot for the association between total fluid intake and risk of bladder cancer . P valueTo our knowledge, the current study represents the most comprehensive and up-to-date review on this topic. In this review and meta-analysis of 26 studies with detailed stratified analyses, we showed that there was no unified association between total fluid consumption and the risk of bladder cancer for all people. Nevertheless, high fluid intake would obviously increase the risk in European male and this relationship might also lie in American residents but was relatively weak. On the contrary however, a protective effect from high fluid intake against bladder cancer was noticed in Asia population, especially in Pakistan. The results indicated that the association between total fluid intake and risk of bladder cancer was largely affected by differences in gender, geography region and even the time when the participants were included. Basing on the findings of the current study and those of other previous related studies, we inferred that it was the source and quality of water as well as diets that much related to bladder cancer risk while volume of fluids consumption played a less important role. Our study also provided the evidence that male would be more easily affected by these environmental and living habit factor than female. To prevent higher risk, according to the evidences found in the current study, it is suggested to limit total daily fluid consumption to 2000ml for European males and 3000ml for America residents. There is a significant relationship that every increment 1000ml daily consumption would increase the risk by 28.6% in European male. Similarly every additional 1000ml consumption may increase the OR by 14.9% in American people but the association is not that strong.What is noteworthy, however, is that the majority of data from Europe and America was collected from cases included before 1990, which may be unable to represent the present conditions. Subgroup analysis by time in the current study indicated that when pooled together, cases recruited before 1990 showed positive outcomes while more recent studies provided negative ones. This difference may mainly generate from the evolution in disinfection techniques and regulations. Taking America for example, since the first regulation on disinfection by-products was passed in 1979, chemical substances including trihalomethanes and free residual chlorine have been successively monitored . Usage oAnother point to be noted is that significant differences were only observed in case-control studies and the outcome varied among regions. Total water intake had a greater impact on residents in European and America than in Asia. Apart from the potential effects of ethnic difference, diet habits, components of daily fluid intake more specifically, might also played important roles. The Chinses are used to drink boiled water, which may further eliminate organic residues in water and reduce the harm taken by them. Furthermore, as one of the most popular beverages in Asia, tea is proved to have protective effect on several cancers, bladder cancer included . PolypheRecently, Buendia et al.  observedAltieri et al.  and Bai The limitations of this meta-analysis should be acknowledged. First, only published data were included, making publication bias particularly inevitable even though no significant evidence was observed by statistic analysis. Second, we found that the majority of the data in included studies were collected from cohort or diagnosed cases recruited over 25 years ago, which may unable to reflect the exact conditions of the present. More well-designed studies are warranted to draw more proper conclusions.In conclusion, the current study indicates that high fluid intake seems to be a risk factor for bladder cancer in European male and American residents but a protective effect in Asian. Geography region seem to be stronger influence factor on fluid intake and risk of bladder cancer. The reason may lie in differences in water quality and drinking habits. Male are proved to be more susceptible to these potential risk factors. For protective purpose, a recommended total fluid consumption is listed as followed: limitation to 3000ml per day for people living in America and <2000ml per day for European males are recommended. Drinking more boiled water and tea may also help. The actual recommended volume may be higher than mentioned above because a large part of the included data were collected in the past time when water quality wasn't managed as well as the present. So more new case-control and cohort studies are needed to draw more proper conclusion.The study was designed under the guidance of the Meta-Analysis of Observational Studies in Epidemiology guidelines for study reporting .We searched the Web of Science, PubMed and EMBASE to identify articles published up to June 2016. Search terms included “fluid or water” and “intake or consumption” and “bladder cancer or urothelial cancer or transitional or bladder neoplasm or bladder carcinoma”. No language limitations were imposed.Inclusion criteria were as followed: (1) case-control or cohort study exploring the relationship between total fluid consumption and bladder cancer risk, (2) provided exact data in both case and control groups (participants for cohort studies) and (3) adjusted effect estimates with their 95% CIs were directly given or could be calculated from the present data.Studies with overlapping or insufficient data were excluded.Two authors (LQ and TY) performed the selection work. Screening was done by titles and abstracts reading. Full text were reviewed when the abstract was insufficient to determine if the study met the inclusion or exclusion criteria. The references of all studies included were manually searched to identify additional studies. Final agreement on inclusion was made by consensus with all authors. All data, including evaluation of study characteristics, risk of bias, and outcome measures, were extracted by two authors.The Newcastle-Ottawa Scale (NOS)  was used2 score calculating [2 >50% was considered as presence of heterogeneity. In cases lacking of heterogeneity, the Mantel-Haenszel fixed-effect model was used to provide summary estimations, otherwise, the DerSimonian and Laird random-effect model was adopted for the meta-analysis [The ORs were extracted and transformed to their natural logs for each include study. The log ORs were weighted by the inverse of their variances to obtain a pooled OR with 95% CI. The Q-test was adopted for statistical heterogeneity measured  and the culating . Heteroganalysis , 48. A dPublication bias was assessed through funnel plots as well as tests of Begg rank correlation and Egger regression asymmetry. P<0.05 was considered to be representative of a significant statistical publication bias. Subgroup analyses were conducted as followed to explore the potential heterogeneity among studies: gender , study design (cohort and case-control studies), geography region , cases recruited year (before 1990 and after 1990), Fluid intake quantity (selected based on the characteristics of included studies).Statistical significance was set at P<0.05. STATA version 12.0  and Review Manager 5.2  was used for the statistical analyses.


NFKB) is associated with the pathogenesis of numerous malignancies. According to several studies, NFKB1 −94ins/del ATTG promoter polymorphism is associated with the risk of different malignancies, but the results were not consistent. Therefore, we performed an updated meta-analysis based on 37 case-control studies from 33 articles  to clarify the relationship. The odds ratio (OR) and 95% confidence interval (CI) were used to determine the strength of the association. We found that the NFKB1 −94ins/del ATTG promoter polymorphism was significantly associated with increased susceptibility to cancer in the recessive , homozygote , and allele  genetic models. The subgroup analysis for ethnicity found that the NFKB1 −94ins/del ATTG promoter polymorphism was significantly associated with an increased susceptibility to cancer in Asians and with a decreased susceptibility in Caucasians. The stratified analyses revealed significant associations between the polymorphism and increased susceptibility to ovarian cancer, oral squamous cell carcinoma, and nasopharyngeal carcinoma.As a cell survival signal, nuclear factor-kappa B ( NFKB1 −94ins/del ATTG promoter polymorphism.Cancer is the result of complex interactions between inherited and environmental factors, which threatens people worldwide due to high morbidity and mortality . AlthougNFKB is responsible for regulating the expression of many genes for immune response, cell adhesion, differentiation, proliferation, angiogenesis and apoptosis [NFKB was first identified by Sen and Baltimore in 1986 [NFKB binds to a 10 bp DNA element in kappa immunoglobulin light-chain enhancer in B cells [NFKB family consists of p50/p105, p65/Rel A, c-Rel, Rel B, and p52/p100. Among them, the major form of NFKB is a heterodimer of the p50/p105 and p65/Rel A subunits that are encoded by the NFKB1 and NFKB2 genes, respectively [NFKB1 gene, located on chromosome 4q24, encodes a 50 kDa DNA-binding protein that can act as a master regulator of inflammation and cancer development [poptosis . NFKB wa in 1986 . As a tr B cells . The NFKectively . The humelopment .NFKB1 gene elicits a regulatory effect on the NFKB1 gene [NFKB1 −94ins/del ATTG promoter polymorphism and cancer risk [NFKB1 −94ins/del ATTG promoter polymorphism and cancer, but the results were inconsistent. Since then, several other studies [NFKB1 −94ins/del ATTG promoter polymorphism and susceptibility to a variety of cancers. Therefore, to better understand the precise relationships, we performed a comprehensive updated meta-analysis with increased statistical power.A common insertion/deletion polymorphism in the promoter region of the KB1 gene  and an icer risk . Howevercer risk –48 focus studies –44 perfoOur electronic database search resulted in 202 articles and 2 articles were available manually, we scanned all of the abstracts, and there were 45 articles that conformed to the inclusion criteria, we excluded 9 articles –60 that NFKB1 −94ins/del ATTG promoter polymorphism and cancer risk are shown in Table P values displayed obvious heterogeneity between the selected research studies under all five genetic models of the updated meta-analysis. Thus, the random-effect model was used. We found that the NFKB1 −94ins/del ATTG promoter polymorphism was significantly increased cancer risk in homozygote , recessive  and allele  genetic models. However, the association was not found in II+ID vs. DD  and ID vs. DD .  and decreases in susceptibility in three models among Caucasians . , oral squamous cell carcinoma  and nasopharyngeal carcinoma  in the models. However, we did not find associations in hepatocellular carcinoma, colorectal cancer, bladder cancer, prostate cancer, non-small cell lung cancer and renal cell cancer  in this updated meta-analysis.The publication bias analysis was performed by Begg's funnel plot and Egger's test. The shape of the Begg's funnel plots seemed symmetrical Figure  and EggeThe sensitivity analysis was performed by the sequential omission of individual studies. After excluding each study sequentially, we obtained statistically similar results (data not shown), suggesting that the data of our meta-analysis are relatively stable and credible. In addition, the random-effects model was compared with the fixed-effects model, and the statistically similar results were obtained in all genetic models.NFKB1 −94ins/del ATTG promoter polymorphism and risk of cancers [In recent years, several investigators reported the association between the  cancers –35 such  cancers –48 had t cancers –44, inclNFKB1 −94ins/del ATTG promoter polymorphism and cancer susceptibility. We found that the NFKB1 −94ins/del ATTG promoter polymorphism was significantly associated with increased risk of cancer; this result was different than a previous meta-analysis [NFKB1 −94ins/del ATTG promoter polymorphism and cancer risk. The reasons for this difference could be explained as follows: 1) we included 37 case-control studies, versus only 11 studies  in the previous meta-analysis, and therefore, the results of this meta-analysis were more credible; and 2) there may be some factors among the study populations that could influence the results, including age, gender, life style, and environment. In addition, when compared with the meta-analysis by Wenyuan Duan [NFKB1 −94ins/del ATTG promoter polymorphism with increased and decreased cancer risk in Asian and Caucasian populations, respectively. Surprisingly, the results were different from the result shown by a previous meta-analysis [NFKB1 −94ins/del ATTG promoter polymorphism was associated with risk in Asians but not in Caucasians population. The results may be explained by the following: 1) this discrepancy may be because of the limited sample size. The previous meta-analysis included only 9 articles (2047 cases and 2040 controls) in Caucasians, whereas we included 11 articles (3069 cases and 6584 controls); 2) we excluded the studies that do not follow HWE. Therefore, the results of this study are more reliable than the previous meta-analysis; 3) The sensitivity analysis was conducted through two methods in this meta-analysis, and the results were consistent with the previous results, suggesting the results of this study were stable.We performed a meta-analysis of 37 case-control studies from 33 articles  to clarify the relationship between the analysis , which ruan Duan , althouganalysis , which cNFKB1 −94ins/del ATTG promoter polymorphism was just one of susceptibility genes, and all these non-genetic factors could influence the expression of the gene. Therefore, the differences in this NFKB1 polymorphism in Asians and Caucasians may result from different genetic background, environment, lifestyle or other factors.Although the mechanism was not clear, we assumed that the mechanism underlying the cancer risk was related to the levels of p50. In recent studies ,68, it wNFKB1 −94ins/del ATTG promoter polymorphism and subtypes of cancer, the NFKB1 −94ins/del ATTG promoter polymorphism is a risk factor for oral squamous cell carcinoma, ovarian cancer and nasopharyngeal carcinoma. This result suggests that the NFKB1 gene might have some relevance in these cancers. The inconsistent may be caused by their different micro-environment, because the same genetic factor might have different correlations in different cancer site [According to the results of the analysis of the relationship between the cer site . Our stuThis study has several limitations, like any meta-analysis. First, moderate heterogeneity was detected in some comparisons and may distort the meta-analysis. Second, the non-genetic risk factors such as environment are also important in the incidence ratio of cancer. Unfortunately, there were not enough data for further subgroup analysis; therefore, the results of subgroup analysis may affect the validity of the conclusions. Third, in the subgroup analysis, we found that our analysis was limited to Asian and Caucasian populations, so we do not know whether these conclusions can also be adopted in other populations. This may cause publication bias. Finally, the sample sizes for each type of cancer were relatively small, so further research should enlarge the sample sizes to obtain more accurate conclusions.Despite these limitations, our study has several strengths. First, all of the studies that we chose agreed with HWE, which may increase the validity of the conclusions. Second, the sample size of our study was more than double that of the previous meta-analysis, significantly increasing the statistical power. Although this updated meta-analysis had the above-mentioned shortcomings, we tried to control them through perfected searching, sifting the good ones from the bad and performing the statistical analyses strictly.NFKB1 −94ins/del ATTG promoter polymorphism is associated with cancer risk not only in Asian populations, but also in Caucasian populations. Moreover, there might be a significant association with increased susceptibility between the NFKB1 −94ins/del ATTG promoter polymorphism and ovarian cancer, oral squamous cell carcinoma, and nasopharyngeal carcinoma. Well-designed studies with larger representative sample sizes are necessary to confirm our results.We conclude that the The systematic review and meta-analysis was in accordance with the PRISMA  guidelinesA systematic search of the PubMed, Web of Science, Science Direct, Ovid, China National Knowledge Infrastructure (CNKI) and Wan fang Data electronic databases was performed with the following key words: (“polymorphisms” OR “polymorphism” OR “SNP” OR “single nucleotide polymorphism” OR “variant” OR “mutation”) AND (“neoplasm” OR “cancer” OR “tumor” OR “carcinoma” OR “carcinogenesis”) AND (“NF-κB1” OR “Nuclear factor-κB1” OR “Nuclear factor κB1” OR “NFKB1” OR “nuclear factor kappa B1” OR “NF kappa B1” OR “nuclear factor kB1” OR “rs28362491”).NFKB1 −94ins/del ATTG promoter polymorphism and cancer risk; 3) studies describing the genotype distributions in detail to calculate the OR and 95%CI in cases and controls; and 4) the distribution data in controls must be consistent with Hardy-Weinberg Equilibrium (HWE).No language or other restrictions were imposed in this study and the inclusion criteria were as follows: 1) case-control design; 2) studies evaluating the association between the The exclusion criteria in this meta-analysis were as follows: 1) not concerned with cancer risk; 2) only a case population; 3) duplication of a previous publication; 4) the control group does not conform to HWE; and 5) animal studies.According to the criteria listed above, information was carefully extracted from eligible studies independently by each investigator (Y.Q.L. and D.W.). The following information was collected from each study: surname of the first author, year of publication, ethnicity of subjects, genotyping method, frequencies of the genotypes in cases and controls, cancer type. The different ethnicities were categorized as Caucasian or Asian. Studies that investigated more than one type of cancer were regarded as individual datasets only in subgroup analyses according to cancer type. Any discrepancy was resolved through discussion.NFKB1 −94ins/del ATTG promoter polymorphism and cancer was estimated through OR with 95% CI. The combined ORs were determined by the Z test, and a P value of <0.05 was considered to be statistically significant. The NFKB1 −94ins/del ATTG promoter polymorphism consists of three genotypes: homozygote insertion or wild-type (II), homozygote deletion or variant (DD), and heterozygous ins/del (ID). We measured the association based on five different genetic models: the dominant (II+ID vs. DD), recessive (II vs. ID + DD), homozygote (II vs. DD), heterozygote (ID vs. DD), and allele (I vs. D) models. To investigate the origin of heterogeneity, subgroup analyses based on ethnicity (Caucasian and Asian) and cancer type were performed to identify the association between the NFKB1 −94ins/del ATTG promoter polymorphism and cancer susceptibility.The strength of association between the Q and 2I statistical tests to check the statistical heterogeneity among studies. If the P value was < 0.05 and 2I ≥ 50% indicating heterogeneity, then a random-effect model was chosen to calculate the pooled OR; otherwise, a fixed-effect model was selected [p<0.05 and the plot was asymmetrical [We used the selected . A sensimetrical . The sta


Origin of anorectal malformations (ARM) are considered multifactorial. Several genetic and non-genetic risk factors are discussed in literature. Maternal periconceptional medical drug use as possible risk factor, however, has not been reviewed systematically.Studies published between 1977 and April 2017 were reviewed through systematic search in PubMed, ISI Web of Knowledge and Scopus databases. Furthermore, related and cross-referencing publications were reviewed. Pooled odds ratios  were determined to quantify associations of maternal periconceptional use of folic acid, multivitamins, anti-asthma medication , thyroid hormone supplements, psychiatric drugs  and aspirin with ARM using meta-analyses.n = 14), Sweden (n = 6), Hungary (n = 5), Germany (n = 3), the Netherlands (n = 3), Denmark (n = 2), France (n = 2), Norway (n = 1) and the UK (n = 1). However, only few of these studies reported on the same risk factors. Studies were heterogeneous with respect to case numbers, period ingestion of medical drug use, control selection and adjustment for covariates. Consistently increased risks were observed for any anti-asthma medication, and hypnotics and benzodiazepine, but not for folic acid, multivitamins, inhaled corticosteroids, salbutamol, thyroid hormone supplements, antidepressants, any SSRI, sertraline, citalopram, fluoxetine, paroxetine and aspirin. In meta-analyses, pooled odds ratios  for any anti-asthma medication, and hypnotics and benzodiazepine were 1.64 (1.22–2.21), and 2.43 (1.03–5.73), respectively.Thirty-seven studies that reported on the association between maternal periconceptional drug intake and infants born with ARM were included in this review. These were conducted in the United States of America (Evidence on maternal drug use before conception and during pregnancy as risk factor for ARM from epidemiological studies is still very limited. Nevertheless, the few available studies indicate any anti-asthma medication, and hypnotics and benzodiazepine to be associated with increased risks. Further, ideally large-scale multicenter and register-based studies are needed to clarify the role of maternal drug intake for the development of ARM. Anorectal malformations (ARM) are rare birth defects concerning anus and rectum. Approximately 1 in 2500 to 1 in 5000 new born babies are affected . DiffereWe conducted a systematic review and meta-analysis of epidemiological studies to summarize current evidence on the relationship between maternal drug intake and ARM, and to identify knowledge deficits that need to be addressed in future research.A literature search was carried out to identify epidemiological studies assessing the association between maternal medical drug intake before conception and during the first trimester of pregnancy and anorectal malformations. Relevant studies published in English were systematically searched in PubMed, ISI Web of Knowledge and Scopus databases by using various combinations of the following terms: , congenital abnormality, congenital abnormalities, birth defect(s), anorectal malformation(s), anorectal atresia, anal atresia, imperforate anus) AND , drugs(s), medication, medicament, medicine, pharmaceutical, dietary supplements, folic acid, (multi-)vitamins, vitamin A, vitamin B, vitamin C, iron). Duplicate articles were deleted. Each title and abstract was checked for relevance. The full text was reviewed if the abstract indicated that the article reported an association between ARM and maternal use of medical drugs. Furthermore, the identified articles were reviewed for related articles and cross-referring publications.Articles were included if they reported on associations of anorectal malformations with maternal medical drug intake. When available, data of ARM infants with isolated anomalies  were preferred to data of ARM infants with multiple defects. Articles were excluded if the reported number of ARM cases was less than two. ARM infants analyzed only in a group with other anomalies like intestinal or tracheo-esophageal atresias were also excluded because of concern that associations of risk factors with these anomalies might be different from associations with ARM. In addition, articles describing medical drug use in animal models were also excluded. Searches were restricted to English-language articles.Two reviewers independently assessed the articles and extracted the following key information in a standardized manner: first author, year, country, study design, characteristics of the study population, period of data acquisition, assessed medical drug(s) and the respective measures of odds ratio or risk ratio (see below), as well as covariates adjusted for in the analysis. Initial disagreements on classifications of study characteristics were resolved by discussion within the team of authors. Such disagreements included the presentation of case and control numbers in some studies as well as the presentation of the maternal age at the time of data acquisition. In such a case, the authors came together and discussed the topics. As one result, footnotes were included in Table Associations between maternal medical drug intake and ARM are presented by odds ratios (OR) and their 95% confidence intervals (CI). Alternatively, reported risk ratios (RR) are shown. In one case, only prevalence values were presented. Unadjusted values were recalculated by the Review Manager Software, version 5.3.5  to validate the results. When measures of associations were not explicitly reported, they were derived from data provided in the articles.2 and I2 statistics. When the number of studies is low or when sample sizes are small, the power of the χ2 test is low. The I2 measure describes the proportion of total variation in effect estimates across studies that is due to heterogeneity rather than sampling error [2 ≤ 25) and the random effects model when study heterogeneity was moderate to high (I2 > 25) [P < 0.1).Meta-analyses were performed for risk factors for which results were available from at least three studies. Heterogeneity was assessed by the χng error . Fixed aI2 > 25) , 46. IndI2 > 25)  and EggeI2 > 25)  (P < 0.1n < 2), nine articles because they referred to results of already selected articles and further three articles because they reported on ARM cases analyzed in a group with other anomalies. Finally, 37 articles were included in the review. Among the included studies, 10 provided data on the association of ARM with maternal periconceptional dietary supplements, including folic acid, iron and (multi-)vitamin use, five to asthma medication, three to thyroid medication, nine to psychiatric drugs, four to painkiller, four to anti-infectives, two to drugs against nausea and vomiting, two to sexual hormones and further two to other medical drugs.In total, 146,491 articles were found . The remaining studies were conducted in Sweden (n = 6), Hungary (n = 5), Germany (n = 3), the Netherlands (n = 3), Denmark (n = 2), France (n = 2), Norway (n = 1) and the UK (n = 1). Recruitment was population−/national-based in 30 studies, province-based in one study and hospital-based in four studies. For data acquisition, two studies relied on register-based data [Details on the 37 studies, which were published from 1977 to April 2017, are shown in Table sed data , 50. Datsed data , 51 to 2sed data .Case numbers ranged from six ARM cases  to 799 AMost of the studies asked for a period ingestion of medical drug use before and during pregnancy. However, there was no unique definition of the “periconceptional” period of time, varying from 3 months before conception until the end of the third month in pregnancy , 61, 66 Study results as well as the covariates adjusted for are shown in Tables Ten studies reported on the association between maternal use of dietary supplements before or during pregnancy and infants born with an anorectal malformation, among them seven studies on folic acid, six studies on multivitamins, and each one study on vitamin E and iron intake Table .crude, 0.4; 95% CI, 0.17–0.88; P = 0.01). In contract, the use of folic acid in the first month of gestation was not significant . In the study by Myers et al. [crude, 0.5; 95% CI, 0.29–0.88) and a marginally protective association with ARM cases with no additional external anomalies . After adjustment for maternal age the association with all ARM cases was weakened  and the association with ARM cases with no additional external anomalies became insignificant .The study by Czeizel et al.  reporteds et al.  differencrude, 1.6; 95% CI, 1.0–2.7; P = 0.09). No other study could confirm an association with ARM. The study by Gilboa et al. [adj, 1.7; 95% CI, 1.01–2.72). There was no association with maternal iron intake before or during pregnancy.A marginally increased risk for multivitamins was only reported in the study by van Rooij et al.  . The estimated heterogeneity variance was tau2 = 0.0226. In meta-analysis, no significant association was observed in pooled analyses using the fixed effects model . There was a weak evidence of publication bias .The result of the meta-analysis on the association between maternal use of folic acid and ARM infants is shown in Fig. 2 statistic indicated low heterogeneity across studies . The estimated heterogeneity variance was tau2 = 0.0523. No significant association was observed in pooled analyses using a fixed effects model . There was no evidence of publication bias .The result of the meta-analysis on the association between maternal use of multivitamins and ARM infants is shown in Fig. adj, 1.7; 95% CI, 1.11–2.56) and Garne et al. [adj, 1.6; 95% CI, 1.08–2.51) and the comparison of ARM with chromosomal anomalies  was significant for any anti-asthma medication. In contrast, the study by Zwink and Choinitzki et al. [adj, 2.1; 95% CI, 1.09–4.12).Five studies reported on the association between maternal use of asthma medication before or during pregnancy and infants born with an anorectal malformation Table . The usee et al. . In the i et al.  could noi et al.  observedcrude, 2.0; 95% CI, 1.10–3.51; P = 0.04 and ARM vs. chromosomal anomalies: ORcrude, 3.3; 95% CI, 1.81–5.98; P = 0.02). Both results remain significant after adjustment for center and maternal age . However, results became insignificant after adjustment for more than those two covariates. The study by Källén and Otterblad Olausson [adj, 1.9; 95% CI, 1.00–3.22).The use of inhaled corticosteroids showed a significant risk for ARM in the study by Garne et al. , indepenOlausson  observedcrude, 1.7; 95% CI, 1.08–2.80; P = 0.03 and ARM vs. chromosomal anomalies: ORcrude, 2.3; 95% CI, 1.39–3.75; P = 0.003). After adjustment for covariates the association with the non-chromosomal anomaly control group became insignificant, the association with the chromosomal anomaly control group was weakened . A suggestive association with ARM was observed for the use of budesonide in the study by Källén and Otterblad Olausson [adj, 1.9; 95% CI, 0.95–3.42).Beta-2-agonists in general were not associated with ARM, whereas inhaled beta-2-agonists showed a significant association with ARM in the study by Garne et al.  . In meta-analysis, a significant association was found for maternal use of any anti-asthma medication before or during pregnancy using a fixed effects model . There was no evidence of publication bias .The result of the meta-analysis on the association between maternal use of any anti-asthma medication and ARM infants is shown in Fig. crude, 8.6; 95% CI, 1.7–40.2; P = 0.005) and the study by Källén et al. [crude, 1.9; 95% CI, 1.00–1.85). In contrast, the study by Zwink and Choinitzki et al. [Three studies reported on the association between maternal use of thyroid medication before or during pregnancy and infants born with an anorectal malformation Table . The stun et al.  on a mari et al.  could noadj, 2.5; 95% CI, 1.06–5.68). In addition, the study by Furu et al. [crude, 1.5; 95% CI, 0.95–2.37; P = 0.08). The result became insignificant after adjustment for covariates . Neither the study by Alwan et al. [adj, 2.5; 95% CI, 1.09–5.57; Louik et al.: ORadj, 4.4; 95% CI, 1.2–16.4). Due to the small sample size, confidence intervals were very wide in both studies. However, there was no increased risk for citalopram.Nine studies reported on the association between maternal use of psychiatric drugs before or during pregnancy and infants born with an anorectal malformation Table . The stuu et al.  observedn et al.  could con et al.  and Louin et al. . Both stThree studies assessed antidepressants in general , 56, 76.adj, 1.9; 95% CI, 1.1–3.3) and in the second-third months of gestation . The study by Bonnot et al. [adj, 6.2; 95% CI, 2.44–15.74). No association could be found in the study by Källén et al. [crude, 1.2; 95% CI, 0.33–3.06). A suggestive association with ARM was observed for the use of anticonvulsants in the study by Källén et al. [crude, 2.9; 95% CI, 0.96–6.86).A marginally increased risk for diazepam was found in the study by Czeizel et al.  in both,t et al.  showed an et al.  for othen et al.  . No significant association was found for maternal use of antidepressants before or during pregnancy using a fixed effects model . There was no evidence of publication bias .The result of the meta-analysis on the association between maternal use of antidepressants and ARM infants is shown in Fig. 2 statistic indicated heterogeneity across studies . The estimated heterogeneity variance was tau2 = 0.0744. No significant association was observed in pooled analyses using the random effects model . There was no evidence of publication bias .The result of the meta-analysis on the association between maternal use of any SSRI and ARM infants is shown in Fig. 2 statistic indicated homogeneity across the three studies . No significant association was found for maternal use of citalopram before or during pregnancy using a fixed effects model . There was no evidence of publication bias .The result of the meta-analysis on the association between maternal use of citalopram and ARM infants is shown in Fig. 2 statistic indicated heterogeneity across studies . The estimated heterogeneity variance was tau2 = 0.3820. In meta-analysis, a weak association was observed in pooled analyses using the random effects model . There was no evidence of publication bias .The result of the meta-analysis on the association between maternal use of hypnotics and benzodiazepine and ARM infants is shown in Fig. adj, 2.3; 95% CI, 0.9–5.6) when using a population control group, including maternal self-reported and medical documented drug use.Four studies reported on the association between maternal use of painkillers before or during pregnancy and infants born with an anorectal malformation Table . Only thAmong the four studies that reported on the association between maternal use of anti-infectives before or during pregnancy and infants born with an anorectal malformation Table , none coadj, 0.8; 95% CI, 0.36–1.66) [Two studies reported on the association between maternal use of drugs against nausea and vomiting before or during pregnancy and infants born with an anorectal malformation Table . The stu36–1.66) .adj, 1.2; 95% CI, 0.6–2.3) or estrogens .Among the two studies that reported on the association between maternal use of sexual hormones before or during pregnancy and infants born with an anorectal malformation Table , neitherAmong the two studies that reported on the association between maternal use of other medical drugs, among them antispasmodics, other miscellaneous medication and mixed preparations, before or during pregnancy and infants born with an anorectal malformation Table , none coThis systematic review and meta-analysis summarized the results of 37 epidemiological studies on the association between maternal medical drug intake and infants born with an anorectal malformation reported between 1977 and April 2017. The majority of the studies were conducted in the United States. Case numbers ranged from six ARM cases in the study by Bonnot et al.  to 799 AThere is a great discrepancy in the reported results on the association between the different maternal medical drugs and ARM which impede comparability. As the exact active agent, dose and frequency of medical drug use is not reported in all studies, one can only speculate about possible detrimental effects on embryogenesis due to high dose/overdose of medical drugs. Such effects have been already observed in previous studies with vitamin A and etretinate for other birth defects , 83. On In contrast, the use of folic acid before and during pregnancy is discussed as having a protective effect for birth defects, among them neutral tube defects, cleft lip and cleft palate and heart defects –86. The Depressions are common in pregnancy. Previous studies have been shown that approximately 10–15% of all women suffer from peripartum depression . It is sThe prevalence of asthma in pregnancy is estimated to be 4–12% , 102. MoDisorders of thyroid function are divided into hypothyroidism with an extremely low level of metabolism hormone thyroxine, and hyperthyroidism with an excessive production of thyroid hormones. In general, changes in thyroid function are available in up to 15% of all pregnancies, with a prevalence of overt and subclinical hypothyroidism of approximately 0.4% and 2–3%, respectively, and of overt and subclinical hyperthyroidism of approximately 0.1–0.4% and 2%, respectively , 105. AnWhen available, data on ARM infants with isolated anomalies  were preferred in this review to data on ARM infants with multiple defects. Only three of the 37 reviewed studies looked at both groups , 59, 73.Looking at some other gastrointestinal malformations, asthma medication use during pregnancy also seem to increase the risk for esophageal atresia, omphalocele and gastroschisis , 73. TheThe significant associations with ARM and some other gastro-intestinal malformations show clearly the need of a physician well-controlled medication intake during pregnancy to early detect a possible overdose or incorrect intake or even interactions due to multiple medication intake.Our review has a number of limitations mostly resulting from the overall scarcity of published evidence. First, our meta-analysis was limited by the data provided in the individual studies. Not all studies provided risk estimates adjusted for potentially influential confounders, such as maternal age, periconceptional smoking, pregnancy BMI, race/ethnicity, education and parity. A homogenous epidemiological study is almost impossible to get. It is essential to adjust for potentially confounders, including interaction between various drugs. Doing not so makes a direct comparison of the results difficult and thus the interpretation in meta-analyses. Furthermore, so called “ad hoc” studies would be desirable. However, in studies assessing rare diseases, high efforts and costs are needed to achieve a suitable sample size. Due to the small number of studies, we decided to pool adjusted and crude values for meta-analyses. Second, the used medical drug was almost not exactly described in the studies, including medication classification with its active agents, dose and frequency of medical drug use, and the intake of multiple medications simultaneously and thereby possible occurring interactions. Third, studies did not differentiate between ARM phenotypes ranking from lower to higher forms with different genetic background . Fourth,To our knowledge, our article is the first systematic review and meta-analysis that provides an overview of the available epidemiological studies that reported on the association between maternal medical drug use before and during pregnancy and ARM. Adequate evidence is still very limited, especially in regard to gene interaction. Separate report of isolated ARM and those cases with multiple defects should become standard. Due to small sample sizes, it is understandable that data are very often analyzed together. Approximately 64% of all ARM patients have one or more additional extra-anal anomalies and only 36% have an isolated ARM (no further major birth defect) . Neverth


Previous epidemiological research suggests polymorphisms in long non-coding RNA (lncRNA) H19 are associated with an increased risk of cancer, but the results are inconsistent. We therefore conducted a meta-analysis to more accurately determine the association between lncRNA H19 polymorphisms and cancer risk. The PubMed, Embase, and Science Citation Index online databases were searched and 11 relevant studies involving a total of 33,209 participants were identified. Odds ratios (ORs) and corresponding 95% confidence interval (CIs) from these studies were used to detect associations between H19 polymorphisms and cancer risk using five genetic models. The pooled result suggested that the rs2839698 G>A polymorphism was associated with digestive cancer risk in all five models. Moreover, a protective effect against cancer development was observed for the T allele variant of the rs2107425 C>T polymorphism, especially in Caucasian patient populations. No significant associations were found between lncRNA H19 rs217727 G>A polymorphism and cancer risk. In summary, the rs2839698 G>A and rs2107425 C>T polymorphisms in lncRNA H19 may therefore play opposing roles during cancer development, and their effects may vary depending on cancer type and patient ethnicity. Cancer is a leading cause of death worldwide and has major economic impacts in both developed and developing countries , 2. In 2Increasing evidence indicates that abnormal gene regulation and expression result in abnormal protein activity and disruption of the cell cycle, in turn leading to tumor formation. Long non-coding RNAs (lncRNAs), which were first identified in the 1990s, are single-stranded, non-coding RNAs with lengths of more than 200 nucleotides and no open reading frames . MoleculThousands of lncRNA molecules have been identified to date in the human genome. The lncRNA H19 is located on chromosome 11p15.5 and is 2.3 kb in length . The H19Genetic mutations, such as single nucleotide polymorphisms (SNPs), also influence susceptibility to cancer . PreviouA systematic literature search identified 932 potentially relevant articles. At the end of the gradual selection process, eleven published articles involving a total of 14,030 cancer patients and 19,179 healthy controls met our inclusion criteria and were included in this meta-analysis   Table . FurtherThe sensitivity analysis, which was conducted by omitting studies one by one to examine the stability of the pooled ORs, revealed a slight change when data from Li et al.'s study was removed  Supplem. CumulatP = 0.77; GA vs. GG: P = 0.62; AA vs. GG: P = 0.52; GA+AA vs. GG: P = 0.62; AA vs. GG+GA: P = 0.60)  .P = 0.86, I2 = 72.6%; GA vs. GG: OR = 0.88, 95% CI= 0.73–1.06, P = 0.18, I2 = 65.0%; AA vs. GG: OR = 0.60, 95% CI = 0.26–1.39, P = 0.23, I2 = 16.0%; GA+AA vs. GG: OR = 0.94, 95% CI = 0.78–1.12, P = 0.47, I2 = 66.3%; AA vs. GG+GA: OR = 1.19, 95% CI = 0.93–1.53, P = 0.16, I2 = 64.9%)  indicated protective effects on cancer risk in population-base subgroup analysis. Additional subgroup analyses based on ethnicity and cancer type did not identify any significant associations. Meta-regression analyses and stratified analyses were also conducted, but no critical factors were found that explain the observed heterogeneities.The association between lncRNA H19 rs217727 G>A polymorphism and cancer risk was examined in five studies involving 3,334 cases and 3,768 controls. No significant overall associations were found in any of the five genetic models  Table . In contP = 0.91; GA vs. GG: P = 0.70; AA vs. GG: P = 0.90; GA+AA vs. GG: P = 0.86; AA vs. GG+GA: P = 0.79)  .P = 0.05, I2 = 70.4%; CT+TT vs. CC: OR = 0.87; 95% CI = 0.78–0.97, P = 0.02). The same association with decreased risk was also observed in Caucasian populations in the subgroup analyses by race .The association between the H19 rs2107425 C>T polymorphism and cancer risk was examined in eight studies involving 10,856 patients and 15,612 healthy controls. A protective effect on cancer development was observed in the variant T allele   .Cancer is currently a main cause of poverty and death worldwide. However, the fundamental mechanism of the pathogenesis of cancer remains unclear. Noncoding RNAs, including lncRNAs, affect many cellular processes by regulating gene function and maintaining cellular homeostasis, both of which are closely related to organ differentiation and disease development , 33. AbnLarge numbers of SNP loci have been identified in many lncRNAs, and an increasing number of studies have focused on the association between lncRNA polymorphisms and cancer risk. LncRNAs are approximately 200 nucleotide long RNA molecules that are stably expressed in plasma but do not code for proteins. Accumulating evidence has demonstrated that abnormal expressions of lncRNAs are often promotes tumorigenesis through by interfering with normal cell cycle progression . The H19Meta-analysis can be used to integrate data from multiple studies, thereby expanding sample sizes and increasing the strength of conclusions . We condTo the best of our knowledge, this is the first meta-analysis to examine the association between these three lncRNA H19 polymorphisms and the risk of different types of cancer. Although only eleven studies involving relatively small sample sizes of 14,030 cases and 19,179 controls were included, these results help to characterize associations between lncRNA H19 polymorphisms and cancer risk. Our conclusions are further supported by the observation that genotype distributions of the selected SNP loci in the healthy controls were consistent with HWE and by the lack of any significant publication bias. However, due to the relatively small number of studies included and to limitations of the linear regression method employed, additional studies should be conducted to confirm these results and reduce the effects of possible publication biases.Some additional limitations of this meta-analysis should be considered when interpreting the results. First, only eleven studies involving the three SNP loci examined were identified; the small number sizes might decrease the reliability of the results and increases the probability of random errors, thus affecting the assessment of associations between these lncRNA H19 polymorphisms and cancer susceptibility. Second, some important patient characteristics, such as cigarette use and alcohol consumption, as well as other environmental factors, were not considered. The impact of interactions between genetic and environmental factors on cancer development could not be assessed in this study. Third, all of the included studies focused almost exclusively on Asian or Caucasian populations, and our results may not be applicable to all populations. Fourth, the lncRNA H19 rs2839698 G>A, rs217727 G>A and rs2107425 C>T polymorphisms were analyzed separately, and the effects of haplotype and gene-gene interactions could not be analyzed with the available data. Fifth, heterogeneity existed in all three polymorphism loci. Clinical diversity , methodological diversity  and statistical heterogeneity were the most common source. In our meta-analysis, many factors such as the diversity of cancer type, classification of disease severity, environment factors, might have increased heterogeneity and influenced our findings.Despite these limitations, our meta-analysis indicated that the lncRNA lncRNA H19 rs2839698 G>A and rs2107425 C>T polymorphisms might play different role during cancer development. More studies in patients of different ethnicities and with larger number sizes are needed to confirm these findings.Two of the authors independently searched the online PubMed, Embase, and Science Citation Index (SCI) English databases for relevant studies published prior to December 1, 2016 that examined associations between H19 lncRNA polymorphisms and cancer risk. The bibliographies of relevant studies and recently reviews were retrospectively examined to identify additional articles. The following search terms and strategy were used:#1 lncRNA H19#2 long non-coding RNA H19#3 H19#4 rs2839698#5 rs2107425#6 rs217727#7 #1 OR #2 OR #3 OR #4 OR #5 OR #6#8 polymorphism#9 variant#10 mutation#11 #8 OR #9 OR #10#12 cancer#13 tumor#14 neoplasm#15 #12 OR #13 OR #14#16 #7 AND #11 AND #15Studies were collected according to the following inclusion criteria: 1) case-control studies focused on the association between lncRNA H19 polymorphisms and cancer risk; 2) genotype distribution and frequency data provided were sufficient to estimate the odds ratios (ORs) and 95% confidence intervals (CIs); 3) studies published only in English; and 4) data for the largest sample set or most recent samples were included when data from the same set of patients were used more than once within a study.Two investigators (Li and Yin) independently extracted the following relevant information from the included studies: abbreviated name of the first author, publishing date, country or region where the study was conducted, patient ethnicity (Asian/Caucasian), the design of the controls, sample sizes for patients and healthy controls, genotype distribution frequency data, genotyping method, Hardy-Weinberg equilibrium (HWE) for the controls, and cancer type. Quality evaluation of included studies was assessed by the first two authors using modified Newcastle-Ottawa scale (NOS), the scores ranged from 0 points (worst) to 10 points (best) Table .Q test and I2 statistic[2 value less than 50%[P value < 0.05 was considered statistically significant.Crude ORs with 95% CIs were calculated to assess the strength between lncRNA H19 rs2839698 G>A, rs217727 G>A and rs2107425 C>T polymorphisms and cancer susceptibility. For the lncRNA H19 rs2839698 G>A polymorphism, the following five genetic models were used: allele contrast (A vs. G), co-dominant models (GA vs. GG and AA vs. GG), dominant model (GA+AA vs. GG), and recessive model (AA vs. GG+GA). Similar genetic models were also used to assess the lncRNA H19 rs217727 G>A and rs2107425 C>T polymorphisms. Stratified measurements were calculated based on ethnicity difference, HWE status, control design, and so on. Heterogeneity was calculated using Cochran's statistic. The fix than 50%, 44. Met than 50%, 46. All


HOX transcript antisense intergenic RNA (HOTAIR) have been recently associated with susceptibility to different cancers. Here, a meta-analysis was performed to derive a more precise estimation of the involvement of HOTAIR polymorphisms in cancer development. Data from cases  and controls  were extracted from eligible studies (n = 10) identified in a comprehensive literature search conducted in PubMed, Embase, and the Web of Science databases through January 20, 2016. Overall, association between polymorphism rs920778 and increased cancer risk was significant in allele contrast (odds ratio (OR) = 1.239, 95% confidence interval (CI) = 1.032 - 1.487) and recessive models . In subgroup analysis based on ethnicity, a significant association between polymorphism rs920778 and cancer susceptibility was observed in Asians under all models, but was most compelling under recessive  and homozygous models . Subgroup analysis by cancer type revealed a significant association between polymorphism rs4759314 and susceptibility to gastric cancer in allele contrast , dominant , and heterozygous models . In conclusion, the results indicated that HOTAIR polymorphism rs920778 was more generally associated with cancer risk, particularly in Asians, while rs4759314 was a risk factor for gastric cancer.Polymorphisms in the  Long non-coding RNAs (lncRNAs) are defined as transcribed RNA molecules that are longer than 200 nucleotides and not translated into proteins . AlthougHOX transcript antisense intergenic RNA (HOTAIR), has an oncogenic role in the development of human cancer in diverse tissues [HOTAIR is a 2158-nucleotide lncRNA transcribed from the antisense strand of the HOXC gene which is located on chromosome 12 [HOTAIR represses transcription, but surprisingly, in trans at the HOXD gene cluster on chromosome 2 [Increasing evidence indicates that one of these molecules,  tissues , 7, 8. Hosome 12 . HOTAIR mosome 2 . The lncmosome 2 , which uHOTAIR occurs in many cancers. It is furthermore associated with poor prognosis, and in experimental models, it promotes tumor progression, invasion, and metastasis [HOTAIR single nucleotide polymorphisms (SNPs) have been investigated as potential cancer susceptibility loci and linked to increased risk for human cancers, such as breast [HOTAIR polymorphisms on cancer susceptibility. The results indicated that HOTAIR polymorphisms are associated with increased cancer risk, but mainly in stratified analysis based on ethnicity and cancer type.Numerous studies have demonstrated that overexpression of tastasis –16. Finas breast –19, esops breast , gastrics breast –24, lungs breast , and cols breast . Howevern = 7) and Turkish descent (n = 3). In addition, the studies covered diverse tumor types: gastric cancer (n = 4) [n = 3) [n = 1) [n = 1) [n = 1) [Our database search yielded 10 studies with a total of 7,772 cases and 9,075 controls that were eligible for our meta-analysis –26. The  (n = 4) –24, brea [n = 3) –19, colo [n = 1) , and eso [n = 1) . QualityHOTAIR SNPs extracted from all eligible studies was 7. Of these, only 2 SNPs, rs4759314 and rs920778, were reported in 5 or more studies and were thus the focus of the meta-analysis. The genotype distributions of HOTAIR rs4759314 and rs920778 SNPs are shown in Table The number of rs4759314 or rs920778 was associated with a general risk for cancer regardless of tissue origin. Analysis including all individuals from all eligible studies revealed that association between the polymorphism rs920778 and increased cancer risk was statistically significant in allele contrast  and recessive genotype models . However, no significant association between the polymorphism rs4759314 and increased susceptibility to cancer in any genotype model was observed. The analysis thus revealed an association between the T allele or the TT genotype for polymorphism rs920778 and increased risk for cancer.The first step in the analysis was therefore to determine whether either HOTAIR polymorphisms rs4759314 or rs920778 were associated with risk for a specific cancer type, stratified analyses were performed on the basis of cancer tissue of origin. In this analysis, association of polymorphism rs4759314 with increased gastric cancer susceptibility was statistically significant under allele contrast , dominant , and heterozygous  models , dominant , and heterozygous models . Therefore, the random-effects model was used to pool the data. In the stratified analysis based on cancer tissue of origin, no significant heterogeneity was found in the gastric cancer group.Significant between-study heterogeneity was observed in allele contrast ; dominant model ; recessive model ; homozygous model ; and heterozygous model  . Therefore, the results of this meta-analysis including all eligible studies were generally and statistically stable.The stability of the results of the meta-analysis was assessed in sensitivity analysis, where the effect of each study on the pooled OR was examined by repeating the meta-analysis after sequentially omitting each study. Through the deletion of only a single study which was based on a Turkish population , an assoP-values from the Egger's test and the Begg's test are listed in Table rs920778. For this polymorphism, publication bias was apparent in the allele contrast model. However, analysis with the trim and fill method demonstrated that the results of our study did not significantly change even after adjusting for the publication bias.Publication bias for each polymorphism in each model was evaluated using both the Begg's and Egger's tests. No obvious asymmetry was observed in any of the Begg's funnel plots indicating that publication bias was generally not a factor influencing the results Figure . All P-vHOTAIR in various cancer types has led to its examination as a candidate molecule for the diagnosis and treatment of the disease. Here, HOTAIR was investigated for potential links to cancer susceptibility. We performed a meta-analysis using data from available cases  and controls  in the literature to date to provide further evidence that HOTAIR polymorphisms rs4759314 and rs920778 are associated with cancer risk. Only polymorphism rs920778 was found to be generally associated with increased risk for cancer, but statistically significant associations for both polymorphisms were revealed when data were stratified based on ethnicity and cancer type.The overexpression of rs4759314 might be related to susceptibility for gastric cancer. A possible mechanism underlying the association of these alleles to increased susceptibility is through the regulation of the expression of HOTAIR itself. Polymorphism rs4759314 resides in an intronic promoter region, which was found to influence the activity of this promoter and expression of HOXC11 gene [rs4759314, increased HOTAIR expression originated from the G (relative to the A) allele in luciferase reporter assays. A similar finding was made for polymorphism rs920778, which resides in a novel HOTAIR intronic enhancer. In this case, increased HOTAIR expression was associated with T allele carriers [HOTAIR.The results of our analysis indicated that the T allele or TT genotype of polymorphism is a potential genetic marker for cancer susceptibility, especially in Asians, and that the G allele in polymorphism C11 gene . For polcarriers . For botHOTAIR polymorphisms and cancer risk as compared to individual studies. However, some limitations of this meta-analysis exist. First, although the analysis was performed with strict criteria for study inclusion and precise data extraction, significant between-study heterogeneity existed in some comparisons. However, after analysis stratified by cancer type or ethnicity, heterogeneity between the subgroups was significantly reduced. Second, our analysis was limited to individuals of Asian and Turkish descent so that it remains unclear as to whether these results can be generalized to other populations.A sufficient number of cases and controls were pooled from different studies and provided a more accurate estimation of the associations between the HOTAIR polymorphism rs920778 is more generally associated with cancer risk, particularly in Asians, whereas polymorphism rs4759314 may be a risk factor for gastric cancer. However, these polymorphisms require further evaluation in future well-designed studies as potential genetic susceptibility loci in different cancers as well as ethnic populations.In conclusion, our results indicate that HOTAIR polymorphisms with cancer risk up to the date of January 20, 2016. The literature search was performed using free-text words combined with Medical Subject Headings (MeSH), such as “Neoplasms”, “HOTAIR long untranslated RNA, human” and “Polymorphism, Single Nucleotide”. Gene-specific terms (HOX transcript antisense intergenic RNA or HOTAIR) were combined with polymorphism-specific terms (polymorphism or polymorphisms or variation or variations or variant or variants or mutation or mutations or genotype or genotypes) and disease-specific terms (cancer or cancers or tumor or tumors or neoplasm or neoplasms) to retrieve eligible studies. References cited in retrieved articles were also reviewed to identify additional potentially relevant studies.PubMed, Embase, and Web of Science databases were searched for studies reporting association of HOTAIR polymorphisms and all types of cancer; (3) providing sufficient data for SNP allele and genotype frequencies; (4) published in English; and (5) performed on humans. Exclusion criteria were the following: (1) reviews and comments; (2) performed on animals, and (3) duplication of a previous publication.Inclusion criteria for studies were the following: (1) case-control or cohort study design; (2) evaluating associations between Two investigators (QQ and JW) independently extracted the following data from each study: first author's surname, publication year, ethnicity, cancer types, numbers of cases and controls, and genotype distributions of cases and controls. Different ethnicity descents were categorized as Asian or Turkish. Study design was stratified into population-based and hospital-based studies. The results were compared and disagreement was resolved by discussion with a third reviewer (BH) until consensus was reached.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp; maximum score = 9 points) was used to evaluate the methodological quality, which scored studies based on the selection of patients, the comparability of the groups, and the quality of the sampling process. A study awarded a score of 0 - 3, 4 - 6, or 7 - 9 was considered as a low-, moderate-, or high-quality study, respectively.The Newcastle-Ottawa Scale and Agency for Healthcare Research and Quality  with 95% confidence intervals (CI) was used to estimate the strength of associations, and the significance of ORs was determined with the Z test. Heterogeneity among the included studies was assessed by Chi square-based Q statistic. A random-effects (DerSimonian-Laird method) or fixed-effect (Mantel-Haenszel method) model was used to calculate pooled effect estimates in the presence (P < 0.05) or absence (P > 0.05) of heterogeneity. The method of subgroup analysis according to the ethnicity or cancer type of the participants was also applied to gain more precise results. Sensitivity analysis was conducted by sequentially excluding each study. The Begg's test and the Egger's test were performed to analyze for the presence of publication bias. If any possible bias was observed, the trim and fill method was used to identify and adjust for those studies. Data analyses were carried out using Stata software, version 11.0 . P-values < 0.05 were considered statistically significant.To obtain a more comprehensive assessment of associations between


Findings from this dose-response meta-analysis suggest that Hcy levels are linearly and positively associated with risk of all-cause mortality.Plasma homocysteine (Hcy) levels may be associated with all-cause mortality risk. However, the results of this association are conflicting and the dose-response relationship between them has not been clearly defined. In this meta-analysis, we conducted a systematic literature search of the PubMed, Embase, Web of Science and Cochrane Library for the relevant articles dated up to February 2017. Pooled relative risks (RRs) and corresponding 95% confidence intervals (CIs) were calculated to evaluate the estimates, and the dose-response relationship was estimated using a restricted cubic spline model. Eleven prospective studies  were included. The summary RR of all-cause mortality for the highest Hcy category vs. the lowest Hcy category was 1.80  with the random effects model. In dose-response meta-analysis, Hcy levels were significantly associated with all-cause mortality risk in a linear fashion ( A high level of homocysteine in the blood (hyperhomocysteinaemia) is correlated with a wide range of diseases, including cardiovascular disease, renal dysfunction, fractures, and cognitive impairment5.Homocysteine (Hcy), a sulphur-containing amino acid, is biosynthesized from methionine via a multi-step process8. For example, Vollset SE et al. reported that a 5 µmol/L increment of Hcy was correlated with an increase in all-cause mortality of 49%9. However, during an 11-year follow-up, Swart KM et al. found that elevated plasma Hcy levels were not related to the risk of mortality in older men10. Recently, a meta-analysis8 with six studies showed that elevated Hcy levels increased the risk of all-cause mortality in women  and in the whole population , but not in men . Moreover, this meta-analysis with three studies found that the pooled RR of all-cause mortality was 1.27 for a 5-μmol/L increment in serum Hcy in the whole population, implying a possible potential dose-response. However, the small number of studies limits the statistical power and a further dose-response meta-analysis. Therefore, considering the above mentioned findings, it is necessary to assess the association between Hcy levels and risk of all-cause mortality with a larger dataset using a dose-response meta-analysis.Additionally, the association between Hcy levels and risk of all-cause mortality in human subjects has been explored, but the results are conflicting. In several follow-up studies, Hcy levels were reported to be associated with all-cause mortality18 with a total of 27,737 participants  were included. Of the included studies, 3 studies were performed in Asia, 2 in North America, and 6 in Europe. Sample sizes ranged from 215 to 7166, and the median follow-up time varied from 3.08 years to 11 years. Table The study selection process is presented in Fig. I2 = 64.8%, pH = 0.002, pz < 0.001) , publication bias might exist, although the Begg’s test was not statistically significant (p = 0.119). Because of this, we conducted a trim and fill analysis that conservatively imputes hypothetical negative unpublished studies to mirror the positive studies that cause funnel plot asymmetry. The imputed studies showed a symmetrical funnel plot .Our meta-analysis showed that elevated Hcy levels were associated with all-cause mortality risk. Further, the pooled relative risk (RR)  was 1.80 1.51, 2.14) under the random-effects model with substantial heterogeneity among studies (.14 under01) Fig.  and Eggelot Fig.  and the 16 were included in the dose-response meta-analysis. The result showed a linear relationship between Hcy levels and all-cause mortality (pnon-linearity = 0.255), by using a restricted cubic spline model. Furthermore, the linear model revealed an increased all-cause mortality risk of 33.6% for each 5 µmol/L increase in Hcy levels  . For studies that adjusted for alcohol consumption, body mass index (BMI), physical activity, cardiovascular disease (CVD), and serum vitamin B12/ folate, the pooled RRs (95% CIs) were 1.91 (1.30–2.80), 1.91 (1.53–2.38), 2.10 (1.76–2.50), 1.94 (1.42–2.65), and 1.91 (1.30–2.80), respectively, which were stronger than for those that did not adjust for these factors .In subgroup analyses of the association between Hcy levels and all-cause mortality, all subgroups revealed significant positive associations, except for the studiesSensitivity analysis revealed that no single study significantly changed the pooled results, which indicated that the results of our meta-analysis were robust and reliable Fig. .Figure 5It is important to explore whether Hcy levels were correlated with all-cause mortality. In the current meta-analysis, we identified that elevated Hcy levels were associated with the risk of all-cause mortality. Furthermore, the dose-response meta-analysis showed a linear association between Hcy levels and all-cause mortality risk, and each 5 µmol/L increment of Hcy corresponded to a 33.6% increase in risk of all-cause mortality.8. Although the final conclusion was not significantly changed, there were some differences when comparing the two meta-analyses. First, by searching four databases , the present meta-analysis included more studies . Second, although dose response analysis was conducted in both studies, the methods and results were different. In the previous study, the dose response analysis was conducted based on 3 original articles that reported per 5 μmol/L increment in serum Hcy levels. In our study, the restricted cubic spline model was performed to estimate the dose-response relationship. Additionally, we found that Hcy levels were significantly associated with all-cause mortality risk in a linear fashion; this was not reported in the previous study. Third, compared with the previous study, we conducted adequate subgroup analyses, further investigated the potential inter-study heterogeneity, examined the robustness of primary results, and explored the influence of relative factors reported in previous studies.Previously, a categorical meta-analysis, by including six studies, found that elevated Hcy levels were associated with all-cause mortality risk, and it found a significant dose-response association between them by including three studieset al., in a study including 4,766 individuals, reported that a 5-µmol/L increment of Hcy was correlated with a 49% increase in all-cause mortality, and, by using generalized additive logistic regression9, the increase in mortality was almost linear. Moreover, using restricted cubic spline, Wong YY et al. observed a graded association between Hcy and all-cause mortality17. Similarly, with 27,737 individuals included, the present study observed that elevated Hcy levels increased the risk of all-cause mortality in a linear fashion; this further indicated the association between Hcy levels and all-cause mortality. The potential biological mechanisms leading to the positive association between Hcy levels and risk of all-cause mortality are not well elucidated. It is likely that elevated Hcy levels may increase all-cause mortality risk by increasing the risk of major chronic diseases, including cardiovascular disease19, fracture20, cognitive decline21, and endothelial dysfunction22. Higher Hcy levels may cause platelet activation and thrombus formation23, imbalance between osteoblasts and osteoclasts24, hippocampal atrophy25, and endothelial dysfunction22, leading to cardiovascular disease, cognitive decline, fracture, endothelial dysfunction, which may eventually contribute to mortality. Further studies are needed to indicate the underlying biological mechanisms.In addition, Vollset SE et al.10 reported that elevated Hcy levels increased mortality risk in women only. A previous meta-analysis8 included two studies and identified that elevated Hcy levels increased the risk of all-cause mortality in women ; however, this positive association was not statistically significant in men . Although we included one more study than the previous meta-analysis, we also did not observe the relationship between Hcy levels and mortality risk in men. Given that most previous studies on higher Hcy levels with mortality risk did not study sex differences, the number of included studies was still limited in the present meta-analysis (only three studies were included), and the results might not be robust enough to draw a conclusion. Thus, it is still unclear whether women or men with elevated Hcy had more mortality risk. Ongoing studies are needed.Hcy levels are typically higher in men than in women. Accordingly, several studies revealed that the association between Hcy levels and all-cause mortality might differ by sex. Swart KM 26. Indeed, numerous studies have indicated that elevated Hcy levels were correlated with low levels of vitamin B and folate28. In the present study, we identified that the pooled relative risk for the studies that adjusted for serum vitamin B12/ folate was 1.91 (95% CI: 1.30–2.80); this was stronger than that for studies that did not adjust for this factor , suggesting that vitamin B/folate might reduce Hcy levels and thus decrease the risk of all-cause mortality. This subgroup analysis further demonstrated that increased Hcy levels were associated with all-cause mortality.Plasma Hcy levels are influenced by life-style and sociodemographic factors. For example, vitamin B and folate, as important cofactors, play an important role in Hcy metabolism. Inadequate dietary supply of vitamin B and folate may lead to hyperhomocysteinaemiaAlthough the present study revealed a positive association between Hcy levels and all-cause mortality, the findings should be considered with caution because of the moderate heterogeneity. There are several potential explanations for the observed inter-study heterogeneity. First, with subgroup analyses, we discovered that geographical location, duration of follow-up, sample size, fasting status, and NOS score contributed more or less to the potential heterogeneity, even though no specific factors were found. Second, lack of consistency in the categories of Hcy levels could also have contributed to inter-study differences in the strength of the observed associations. Third, the differences in adjusted confounding factors might have contributed to the heterogeneity to some extent.A strength of this meta analysis was the prospective design (cohort studies and nested case-control studies were included), which should have greatly reduced the potential of selection bias. Numerous cohort studies with relatively large sample size and long follow-up enabled us to perform informative analysis in various subgroups of populations. In addition, the dose-response meta-analysis further provided a comprehensive description of the association of Hcy levels with all-cause mortality risk. Moreover, in this study, we conducted sensitivity analysis to reflect the influence of each study on the pooled estimates by removing one study at a time. We observed that the estimates did not change significantly, suggesting that the result of this meta-analysis was robust and reliable.Despite our important findings, there were several limitations to our study. First, selection bias might have been introduced because of our relatively strict inclusion and exclusion criteria and failure to acquire complete data either from the original manuscript or from the authors; however, little evidence for such bias was observed. Second, the subjects in most of the studies were elderly; thus, whether the positive association between Hcy levels and all-cause mortality risk could be generalized to younger persons needs to be confirmed. Third, lack of consistency in the categories of Hcy levels might have contributed to inter-study differences that affected the strength of the identified associations. Fourth, the significant heterogeneity reduced the credibility of our results to an extent that the findings in current studies should be considered with caution. Fifth, publication bias was found according to the funnel plot and Egger’s test, although the trim and fill analysis did not change the general result (though the strength of the association was slightly attenuated), suggesting that the association is not an artefact of unpublished negative studies.In summary, findings from this dose-response meta-analysis suggest elevated plasma Hcy levels are associated with linearly increasing risk of all-cause mortality. More in-depth studies are warranted to explore the roles of the biological mechanisms and public health interventions aimed at reducing Hcy levels.The current meta-analysis was performed based on the published criteria of the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA).A computerized literature search was performed in PubMed, Embase, Web of Science, and Cochrane Library (dated up to February 2017) to collect articles regarding the correlation between Hcy levels and all-cause mortality. The following search terms were used: “Hcy” or “homocystine”, or “homocysteine”, or “L-Isomer homocysteine” or “2-amino-4-mercaptobutyric acid”, and “mortality” or “death”. Meanwhile, reference lists of relevant articles were also collected.The inclusion criteria were as follows: (1) prospective studies including cohort studies and nested case-control studies that explored the association of Hcy levels with all-cause mortality; (2) had reported estimates of RRs with 95% CIs or had presented data with which to calculate these; (3) the publication with longer follow-up and more applicable information was selected when multiple publications on the same study population were available; and (4) articles published in the English language.The exclusion criteria were as follows: (1) the design was a retrospective study, animal study, meeting abstract, review article, editorial, case report, or other meta-analysis; (2) only unadjusted hazard ratio (HR), RR or odds ratio (OR) were reported, or raw data was unavailable for retrieval; and (3) subjects were in a highly selected disease group.Two reviewers (Rui Fan and Aiping Zhang) independently extracted data, cross-checked the data, discussed all conflicts, and reached a consensus on all items. The following characteristics were extracted from each study: first author, year of publication, study location, design of study, duration of follow-up, sample size, age, sex, Hcy comparison, RRs, HRs or ORs with 95% CIs comparing the highest Hcy category with the lowest Hcy category, and adjustment for covariates. For studies that showed several multivariable-adjusted RRs, HRs or ORs, the risk estimates that adjusted for most confounding variables were selected.29. NOS scores ranged from 0 to 9, and a score of >7 indicated a high quality.The quality of the included studies was assessed independently by the same two reviewers mentioned above, according to the Newcastle–Ottawa Scale (NOS) criteriaCategorical and dose-response meta-analyses were performed to assess the association between Hcy levels and all-cause mortality. Pooled RRs with their 95% CIs were used to evaluate the effect size for the association of Hcy levels with all-cause mortality. For studies that showed results separately for females or males, we pooled the results to obtain an overall estimate before combining the results with other studies.I2 statistic was applied to quantify the inter-study differences; an I2 value of 0% denoted no heterogeneity, 25% denoted low heterogeneity, 50% denoted moderate heterogeneity, and 75% denoted high heterogeneity30. The Q statistic was used to formally test for heterogeneity . The fixed-effects model was employed to calculate the RR when heterogeneity was found; otherwise, the random-effects model was applied. In this meta-analysis, subgroup analyses were further conducted to investigate the potential inter-study heterogeneity and examine the robustness of primary results31. We divided the subgroups as follows: geographical location , sex , duration of follow-up (<7 years17 or ≥7 years18), sample size (<100015 or ≥100018), fasting status (yes15 or no16), and NOS score (≤717 or >718). We also performed analyses stratified by whether the studies had adjusted for potential confounding factors, including smoking, alcohol consumption, BMI, physical activity, history of CVD, history of diabetes, and serum vitamin B12/folate.The 33. A 2-stage random-effects dose response meta-analysis was performed to examine the potential trend between Hcy levels and all-cause mortality35. This was applied by modelling Hcy levels using a restricted cubic spline model with 4 knots (2 spline transformations) chosen at the 5th, 35th, 65th, and 95th percentiles of the exposure distribution36. In the first stage, by taking into account the correlation within each set of specific relative risks, a restricted cubic spline model was fitted to estimate the 2 study-level coefficients and the within-study covariance matrix35. In the second stage, by pooling the study-specific coefficient estimates and variance/covariance matrices that had been obtained in the first stage, we derived the overall estimates with random effects37. To test the hypothesis that the coefficient associated with the second spline was different from 0, a p value for nonlinearity was calculated35.For dose-response meta-analysis, the included studies should have reported the numbers of deaths and persons/person-years for at least 3 Hcy level categories and the mean or median values of the categories, or have presented the estimated midpoints of the categories if these were not shown in the studies. We assumed the open-ended categories were of the same amplitude as the adjacent categories when the highest categories were open-ended38 Egger’s test,39 and funnel plot were conducted to assess the presence of publication bias . The trim and fill analysis was also conducted to further evaluate the possible effect of publication bias in our meta-analysis40. This method considers the possibility of hypothetical “missing” studies that might exist, imputes their RRs, and recalculates a pooled RR that incorporates the hypothetical missing studies as though they actually existed. All analyses were performed using the software STATA version 12.0 . Values of p < 0.05 were considered statistically significant.We performed sensitivity analysis by removing one study at a time to calculate the overall homogeneity. Begg’s test,


Moreover, we also found a statistically significant association between the HO-1(GT)n repeat length polymorphism and RS under allelic  and co-dominant  models. We found a significant association of the HO-1T(−413)A single-nucleotide polymorphism (SNP) with CHD under allelic , recessive , and co-dominant  models. Our study indicates that both the HO-1(GT)n repeat length polymorphism and the T(−413)A SNP are associated with decreased risk of CHD. The (GT)n repeat length polymorphism was associated with RS following PCI.Numerous published studies have suggested that there is association between heme oxygenase-1 (HO-1) gene polymorphisms and coronary heart disease (CHD) or restenosis (RS) after percutaneous coronary intervention (PCI). This study aimed to clarify this association using a meta-analysis method. We used a systematic search for studies on the association of HO-1gene polymorphisms with CHD or RS in PubMed, Web of Science, the Cochrane Library, Wanfang Data and CNKI . We used Stata 12.0 software to perform the meta-analyses. Twenty-three studies, involving 12,130 patients with CHD or RS and 14,181 controls, were included. A statistically significant association between the HO-1(GT)n repeat length polymorphism and CHD was observed under allelic (odds ratio (OR) = 0.929, 95% confidence interval (CI) = 0.881-0.978,  Coronary heart disease (CHD) is a multifactorial disorder resulting from the interaction between environmental and genetic factors . Many geCurrent studies have documented the interaction between HO-1 gene polymorphisms and CHD or RS after PCI. HO-1 is a subtype of heme oxygenase (HO) which plays a key regulatory role in the synthesis and catabolism of bilirubin . HO willAlthough many studies on the relationship between these two loci and CHD have been carried out –20, the Based on these observations, two meta-analyses –28 relatThere were 176 potentially relevant papers acquired from PubMed, Web of Science, the Cochrane Library, Wanfang and CNKI databases. Of these, we excluded 143 documents because of irrelevance to the aim of our study after reading the title and abstract. The remaining 33 documents were full-text reviewed, and 3 studies were excluded due to reported associations with diabetes –31. Four2 test in any of the genetic models n repeat length polymorphism and CHD. No significant heterogeneity was identified by H-test and Ils Table , therefols Table . Howeverls Table .The second subgroup analysis was conducted according to ethnicity. The fixed-effects model was utilized to perform meta-analysis in all of the genetic models. We found patients with SS genotype have decreased CHD risk compared to SL+LL and LL genotype carriers in the Asian subgroup . However, this association was not observed in Caucasian populations  in the good-quality subgroup . However, this association was not found in the poor-quality reports  A SNP and CHD. No significant heterogeneity was found in the contrast models, and so the fixed-effect model was used in this of meta-analysis. Meta-analysis suggested that there was a significant association between the HO-1 T(−413)A polymorphism and CHD under the allele contrast , the recessive genetic model , and the co-dominant genetic model  n repeat length polymorphism and T(−413)A SNP Table .FHIT) , a recessive genetic model , a dominant model , and a co-dominant model . The statistically significant level was determined by Z-test, and significance was set at