BsmI (rs1544410) polymorphism located in intron 8 at the 3′-end of the vitamin D receptor (VDR) gene is known to be involved in the regulation of mRNA stability. Many studies evaluated the possible correlation between VDR BsmI polymorphism and the risk of pulmonary tuberculosis (PTB), and reported conflicting results. In the present study, an updated meta-analysis was performed to evaluate the above-said association. PubMed, Embase, and Google Scholar web-databases were searched for the relevant studies and a meta-analysis was performed by calculating pooled odds ratios (ORs) and 95% confidence intervals (95% CIs) for all the genetic models. A total of 19 studies comprising 3644 controls and 2635 cases were included in the present study. Overall no association of PTB in allelic contrast , homozygous , heterozygous , dominant  and recessive  genetic models in comparison with wild-type allele and genotype BB were observed. However, variant allele  showed increased risk of PTB in Asians. In conclusion, VDR BsmI polymorphism is not a risk factor for PTB in overall population. However, this polymorphism may be interrelated to an increased risk of PTB amongst Asians. Mycobacterium tuberculosis strain itself, the host genetic factors may also determine the differences in the host’s susceptibility toward TB [Pulmonary tuberculosis (PTB) is the most common type of active TB that severely affects the lungs. PTB is still a leading cause of deaths across the world . Regardloward TB . Unfortuoward TB . Previouoward TB . In the 3 acts as a modulator that activates the monocytes and stimulates the body’s immune system [M. tuberculosis in the macrophages [BsmI, which is present on intron 8 bound to the 3′-UTR of VDR gene, is related to the regulation of mRNA stability and half-life [BsmI polymorphism has been shown to be related with VDR activity or expression [BsmI polymorphism may affect gene expression through regulation of mRNA stability [VDR mRNA expression has been correlated with variant genotypes of the VDR gene [Vitamin De system . The actrophages . VDR genrophages . VDR genrophages . Variatialf-life . This poalf-life . BsmI popression ,13. Studtability ,15. AlteVDR gene .BsmI (rs1544410) polymorphism on the susceptibility of PTB in different populations [BsmI genotype is associated with PTB and contributes to higher/reduced risk of PTB. In addition, the individual studies might have been underpowered to detect the overall effects. Most of the published studies are facing the problems of limited sample size, ethnic diversity, and subsequently suffer too low power to detect the actual effects that may actually exist. In view of sufficient amount of accumulated data, we deemed it is logical to perform a quantitative synthesis of the evidence by applying meta-analysis and provide more precise conclusion of the previous findings. Hence, in the present study, a meta-analysis based on a total of 19 independent studies was performed to evaluate the precise association of VDR BsmI gene polymorphism and risk of developing PTB. A meta-analysis is a powerful statistical tool for studying cumulative data from the independent studies, where individual sample sizes are small and bear low statistical power [Keeping the biological significance of this genetic variant in view, several subsequent reports have been published mentioning the impact of VDR ulations –34. The al power . The schA carefully planned literature search was carried out using PubMed (Medline), Embase, and Google Scholar web databases encompassing all the research articles published with a combination of the following keywords: ‘VDR’, ‘Vitamin D receptor’ gene (‘polymorphism’ OR ‘mutation’ OR ‘variant’) AND ‘tuberculosis’ OR ‘TB’ OR ‘pulmonary tuberculosis’ OR ‘PTB’ AND ‘susceptibility’ OR ‘risk’ (last updated on June 2016). All studies that showed potential relevance for genetic association were weighed by examining their titles and abstracts. All the published studies matching with the above-said eligibility criteria were obtained and tested for their eligibility for incorporation in the present meta-analysis.BsmI gene polymorphism and PTB risk, (ii) adopted a case–control study design, (iii) clearly described confirmed PTB patients and TB-free healthy controls, (iv) have available genotype frequency in the ‘cases’ and in the ‘controls’, (v) published in the English language, and (vi) data collection and analysis methodology should be valid from the statistical point of view. Also, when the case–control was reported in more than one research article using the same case series, we selected the only research study that included the largest number of the subjects. The major reasons for the exclusion of the studies from this meta-analysis were: (i) overlapping of the data, (ii) case-only studies, and (iii) review articles.In order to minimize heterogeneity and facilitate the precise elucidation of the results, studies included in the present meta-analysis study had to pass the following criteria: (i) all included studies must have appraised the association between VDR For each retrieved study, the procedural quality assessment and data extraction were independently summarized in duplicate copies by the two independent investigators  following a standard protocol. During the data extraction process, data-collection form were used to ensure the accuracy of the collected data by stringently following the pre-set inclusion/exclusion criteria as mentioned above, and sequential exclusion of the unsuitable studies. In case of disagreement between the above mentioned two investigators on any item related with the data collected from the selected studies, the issue was fully discussed and deliberated with the investigators to reach a final consensus. Also, in case of failure in achieving consensus between the two investigators, the consensus was attained by following an open argument with the adjudicator (S.H.). The main characteristics abstracted from the retrieved publications comprised the name of the first author, the country of origin, publication year, the number of cases and controls, source of cases and controls, study type, genotype frequencies, and association with PTB.The methodological quality evaluation of the selected studies was carried out independently by the two investigators (SAD & RKM) by using the Newcastle–Ottawa Scale (NOS) of quality score assessment . The NOSBsmI gene polymorphism and PTB risk. Heterogeneity belief was scrutinized by the chi-square based Q-test [P-value <0.05) for the Q-test revealed a lack of heterogeneity amongst the selected studies. Fixed/random effects model was used to pool ORs [2I statistics were employed to calculate interstudy variability ranging between 0 and 100%, wherein, 0% indicated no observed heterogeneity and larger values indicate an increasing degree of heterogeneity [ttest  [http://meta-analysis.com/pages/comparisons.html.Pooled odds ratios (ORs) and their respective 95% confidence intervals (95% CIs) were calculated to assess the relation between the VDR d Q-test . The sigpool ORs ,40. Addiogeneity . The Hare <0.05) . The comVDR mRNA or protein expression and review articles were excluded from this pooled analysis. A very stringent criterion was followed in articles’ search, only case–control or cohort design studies with the frequencies of all the three genotypes were included. After thorough analysis and following the stern inclusion and exclusion criteria, 19 originally published studies were found eligible and included in the present pooled study , Google Scholar, and Embase online web-databases. All the retrieved hits/articles were scrutinized by reading their titles, abstracts, and the full texts for the possibly relevant publications. Additionally, the articles were further checked for their suitability for the present meta-analysis , homozygous (bb compared with BB), heterozygous (bB compared with BB), recessive (bb + Bb compared with BB), and dominant (bb compared with BB + Bb), we applied random-effects model to analyze the combined OR and 95% CI for all the comparisons , homozygous , heterozygous , dominant  and recessive  genetic models did not show any risk of the occurrence of PTB in response to the VDR BsmI gene polymorphism as compared with the wild-type homozygous BB genotype  genetic model , heterozygous , dominant  and recessive  did not show any risk of PTB associated with VDR BsmI polymorphism  does not act as a primary susceptibility polymorphism and may be interacting with other causative germ-line polymorphisms found in LD and inhibits VDR functions. However, the present meta-analysis clearly demonstrates that the contribution of VDR BsmI polymorphism to PTB is perhaps very low. Previously, Rashedi et al. [BsmI gene polymorphism in TB patients. Previous report of Wu et al. [BsmI gene polymorphism is associated with a significant decreased TB risk, especially in Asian population, which is inconsistent with several previous independent studies. Also, they found some contradictory results of increased TB risk due to bb genotype in Iranian population and suggested that this may derive from different experimental designs or methods, and warranted for further investigation. In comparison with previously published reports, the present study has major improvements, as it has included only specific PTB cases of relevant published studies. When we studied Asian population separately, we found strong evidence that variant allele b confers susceptibility to PTB in Asians. This finding may help to explain the individual differences in the susceptibility to PTB. A study by Gao et al. [BsmI polymorphism is associated with TB risk only in Asian population. However, more experimental studies with larger sample size or alternative methods must be applied for further investigation to verify such findings as only the mutant allele showed significant outcome.In the present study, associations for the allele contrast, dominant and recessive models were evaluated, and found no evidence of PTB development risk against VDR i et al.  also repu et al.  reportedo et al.  reportedAs it is established that TB is a complex, multifactorial disease influenced by both environmental and genetic factors , hence, Despite the outcomes attained from this meta-analysis and prior to reaching a final conclusion, limitations of this meta-analysis should also be acknowledged. First, we found significant heterogeneity in the overall analysis. Many factors might contribute to this heterogeneity, variation in patients’ characteristics might be an important source of heterogeneity. Some studies used matched controls (e.g. age and sex matched), while, other studies did not perform matching. Second, the reports published only in the English language were considered in the present study. Third and the most important limitation is that the studies indexed by the selected electronic web-databases  were searched for this pooled data analysis. The possibilities are there that some pertinent articles might have been published in other languages and/or indexed in other databases (which are not known to us), may have been missed. The fourth limitation, since the relevant complete data are not available most of the time, hence, we failed to adjust the confounding factors, such as, age, sex, HIV status, and TB severity in this meta-analysis. The fifth constraint was that we were unsuccessful in computing the gene and environmental interactions because of lack of sufficient information in the primary studies.BsmI polymorphism is not associated with risk of PTB. Due to limited database selection, each pooled analysis misses some of the relevant studies. The research study available in one database is not necessarily present in the other one. Hence, each meta–analysis faces limitation of database selection. Likewise, in the present meta–analysis, most reliable databases were considered and some of the missing studies which were not included in other published studies [Despite the above-mentioned drawbacks, there are some strengths of our meta-analysis that support the robustness of the present results. First, this meta-analysis involved a large set of harmonized individual level data from 19 independent studies, which can provide enough statistical power to confirm our results. Second, since funnel plot and Egger’s test indicated no publication bias, thus all the results are statistically robust as per the sensitivity analysis. Also, all the included studies were of good to modest quality fulfilling the preset needful criteria as tested by NOS quality assessment scale. Although, plenty of meta–analyses have been performed in the past considering various case–control studies analyzing this relationship selected from various databases and concluded accordingly –49. For  studies  were incBsmI gene polymorphism with PTB risk and found no evidence of association between VDR BsmI polymorphism and overall PTB risk. Interestingly, we found that variant allele of BsmI polymorphism may contribute to PTB pathogenesis and may help in explaining the individual differences in the susceptibility of PTB in the Asian population. These results suggest that VDR BsmI gene polymorphism is unlikely to be an important risk factor for specific infectious PTB disease. Furthermore, large, well-designed epidemiological studies are needed to understand the roles of VDR BsmI gene polymorphism in the pathogenesis of PTB, especially in the Asian population.In conclusion, this meta-analysis objectively evaluated various independently published studies dealing with the relationship of VDR


Methylation of tumor suppressor gene promoter leads to transcription inactivation and is involved in tumorigenesis. Several studies demonstrate a potential association between the Death-Associated Protein Kinase (DAPK) gene promoter methylation and bladder cancer risk, tumor stage and histological grade. Due to inconsistent results of these studies, we performed this meta-analysis to ascertain the association.Studies were retrieved from the PubMed, Embase, Web of Science and the Cochrane Library databases. Study selection and data extraction were executed by two reviewers independently. Meta-analysis was performed using Stata 13.0 and Review Manager 5.3 software.A total of 21 articles involving 15 case control and 8 case series studies were included in this meta-analysis. DAPK promoter methylation was associated with bladder cancer risk . The frequency of DAPK promoter methylation was equal in bladder cancer tissue and paired adjacent normal tissue . Furthermore, DAPK promoter methylation was associated with higher histological grade  but not associated with tumor stage .The result suggests that DAPK promoter methylation is significantly increased in bladder cancer patients compared to normal controls. DAPK promoter methylation could serve as a biomarker for bladder cancer detection and management. Bladder cancer (BCa) is the 11th most commonly diagnosed cancer in the world . BCa is DNA methylation is involved in transcriptional silencing of tumor suppressor gene, being a common epigenetic event in early phase of tumorigenesis , 8. InacS1 File).This meta-analysis was designed according to the latest version of PRISMA checklist for meta-analysis  and  and (‘‘DAPK or ‘‘Death-associated protein kinase”). References of these publications were manually reviewed in order to retrieve additional studies.Studies were reviewed by two authors (LHD and CM) independently. Studies met the following inclusion criteria were eligible for the meta-analysis: (1) Purpose of study was to evaluate DAPK promoter methylation in either tissue, urine or blood of BCa patients and normal controls or to assess the association between DAPK promoter methylation and tumor stage or histological grade, (2) types of sample from BCa patients and normal controls should be homogeneous, (3) Detection method of DAPK promoter methylation was based on methylation specific polymerase chain reaction (MSP), (4) Articles published in English. When several studies with overlapping data were observed, only those with a larger sample size were included. Studies without sufficient data after contacting with the original author were excluded.Information extracted from each included study were author, year of publication, ethnicity of individuals involved, type of sample, method for methylation examination, pathological stage, histological grade, frequency of DAPK promoter methylation in BCa patients, normal controls and adjacent normal bladder tissue. The normal controls were defined as those without a historical or current diagnose of BCa, including healthy people, those with benign urological diseases and those with tumors of the other kinds. Studies assessing DAPK promoter methylation in both tissue and urine of BCa patients and normal controls were treated as two. Histological grade ≥2 was defined as high grade, otherwise low grade. Tumor stage ≥2 T2 was classified as high stage, otherwise low stage. Data collection was carried out by two reviewers independently (LHD and CM). Different opinions were settled by discussion.Quality of individual study was assessed in accordance with the Newcastle-Ottawa Scale (NOS) assessment separateI2 statistic[I2 values < 50%, the fixed-effects model (the Mantel-Haenszel method) were selected. Otherwise, a random-effects model (the DerSimonian and Laird method) was applied. In addition, a sensitivity analysis was performed to assess the stability of the results. The potential publication bias was examined in a funnel plot visually and the degree of asymmetry was tested by Egger’s test[The pooled odds ratios (ORs) and corresponding 95% confidence intervals (CIs) of different studies were calculated to compare dichotomous statistics between studies. To assess heterogeneity across the studies, Cochrane’s Q test and I2 sstatistic, 26 wereer’s test. This meFig 1. The study by Motlagh et al [S1 Table). Finally, 21 articles [The strategies for study selection and the results were presented in gh et al  examinedarticles , 29–41 wTable 1.The frequency of the DAPK promoter methylation was evaluated in a total of 1,247 samples from BCa patients and 405 from normal controls in 15 case control studies. Among them nine involved Caucasians and six Asians. Six studies applied MSP method and nine studies used qMSP. MethyLight was classified as qMSP method . DAPK prFig 2). We validated the result in stratified analysis by type of sample, ethnicity and detection method (Table 2).Altogether, DAPK hypermethylation was associated with increased risk of BCa  .Then, we investigated the level of DAPK promoter methylation in BCa tissue and adjacent normal bladder tissue. In pooled meta-analysis, the frequency of DAPK promoter methylation was equal in paired BCa tissue and adjacent normal bladder tissue  . On the contrary, DAPK promoter methylation was associated with higher histological grade  .There was no association between DAPK promoter methylation and tumor stage  , indicating that none of the studies dramatically changed the pooled ORs. The results of the other meta-analysis turned out to be reliable and stable after sensitivity analysis  except for studies of the association between DAPK promoter methylation and histological grade . In omitting Friedrich’s [Sensitivity analysis showed that the pooled ORs ranged from 4.25(95%CI: 2.26–8.01) to 5.36 (95%CI: 2.93–9.78)  (data not shown).Funnel plots of all the studies above were listed in DNA methylation plays an important role in epigenetic transcriptional control and genome stability . MethylaPrevious studies demonstrated that the frequency of DAPK promoter methylation was associated with tumor stage , 37 and The DAPK gene is involved in the p53-dependent apoptosis pathway and has been found to be inactivated via promoter methylation in lung cancer , 47. DAPOur study has several limitations. First, most of the studies were retrospective, information and selection biases and the other confounders may not be fully controlled. Second, seven studies were excluded due to insufficient data in evaluating the association between DAPK promoter methylation and tumor stage and histological grade, which may lead to publication bias and influence the reliability of our conclusion. Third, as regard to source of control, we failed to distinguish between the healthy and those with different types of benign urological diseases. Fourth, while only studies published in English were included, studies of high quality in the other languages may be excluded.In conclusion, our meta-analysis demonstrated that DAPK promoter methylation is significantly associated with BCa tumorigenesis. Further well-designed prospective studies with larger sample size may provide more valid evidence of the role of DAPK promoter methylation in BCa tumorigensis and disease management.S1 File(DOC)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S1 FigS1A Fig. Sensitivity analysis from studies of DAPK hypermethylation in matched bladder cancer tissue and adjacent normal tissue. S1B Fig. Sensitivity analysis from studies of association between DAPK hypermethylation and tumor stage. S1C Fig. Sensitivity analysis from studies of association between DAPK hypermethylation and tumor grade.(DOCX)Click here for additional data file.


APEX1 rs1130409 and rs1760944, ATM rs664677, AXIN2 rs2240308, CHRNA3 rs6495309, CHRNA5 rs16969968, CLPTM1L rs402710, CXCR2 rs1126579, CYP1A1 rs4646903, CYP2E1 rs6413432, ERCC1 rs11615, ERCC2 rs13181, FGFR4 rs351855, HYKK rs931794, MIR146A rs2910164, MIR196A2 rs11614913, OGG1 rs1052133, PON1 rs662, REV3L rs462779, SOD2 rs4880, TERT rs2736098, and TP53 rs1042522) showed significant associations with lung cancer susceptibility with strong cumulative epidemiological evidence. No significant associations with lung cancer risk were found for other 150 variants in 98 genes; however, seven variants demonstrated strong cumulative evidence. Our findings provided the most updated summary of genetic risk effects on lung cancer and would help inform future research direction.More than 1000 candidate-gene association studies on genetic susceptibility to lung cancer have been published over the last two decades but with few consensuses for the likely culprits. We conducted a comprehensive review, meta-analysis and evidence strength evaluation of published candidate-gene association studies in lung cancer up to November 1, 2015. The epidemiological credibility of cumulative evidence was assessed using the Venice criteria. A total of 1018 publications with 2910 genetic variants in 754 different genes or chromosomal loci were eligible for inclusion. Main meta-analyses were performed on 246 variants in 138 different genes. Twenty-two variants from 21 genes ( While smoking is the leading cause of lung cancer, genetics plays an important role as less than 20% of smokers develop this deadly disease in their lifetime2 and non-smokers with a family history of cancer have an increased risk of lung cancer3.Lung cancer is the most common cancer and the leading cause of cancer-related mortality around the worldRGS17 as a possible culprit gene6.Genetic variants influencing lung-cancer risk fall into three categories: rare high-risk variants , moderate-risk variants , and common low-risk variants . Family-based linkage studies is most appropriate for high risk variants with high penetrance but more costly to conduct as lung cancer is a common disease and multiple occurrences of lung cancer in a family are less common. To date, the most concrete linkage and fine mapping studies reveal a lung-cancer susceptibility locus at 6q23–25 and 7. Over the past ten years, common genetic variations at 5p15.33 (TERT/CLPTM1L), 6p21.33 (BAT3/MSH5) and 15q25.1 (CHRNA5/CHRNA3/CHRNB4) are identified to modify the lung cancer susceptibility in GWAS13 and GWAS-based meta-analyses15 . However, these only explain less than 10% of the risk contribution to lung cancer16.Based on the “common disease and common variant” hypothesis, genome-wide association studies (GWAS) provide a powerful tool for investigating the genetic association of a complex disease17, breast cancer18, and colorectal cancer19 are a few good examples using systematic meta-analyses with consensus outcomes.Candidate-gene approaches were the mainstay of genetic association studies before the GWAS era. They are relatively cost-effective and easy to perform. Over 1,000 such studies on the lung cancer susceptibility have been published for the past 25 years. However, there are a number of conflicting reports and it is very challenging to find reliable associations from these highly diverse studies. As a method for systematically integrating data from multiple studies to develop a single conclusion with greater statistical power, meta-analysis is a good way to deal with the diverse and fragmented studies. Although some meta-analyses have been performed on lung cancer, most are limited to investigating a single genetic variant, several variants in a gene, or several variants across a pathway. The recent systematic meta-analyses push the limit to all available genetic association studies in a specific disease and help to achieve a comprehensive view to the genetic contributions to the disease. Alzheimer’s disease22. Additionally, for variants with significant associations by meta-analysis, we applied Venice criteria21 proposed by the Human Genome Epidemiology Network (HuGENet) to assess the epidemiological credibility of cumulative epidemiological evidence of these associations, so as to obtain more reliable results. Moreover, to get a better insight of the differences in genetic variations among populations with different characters, associations stratified by ethnicity, histological types, and smoking status were also examined.Establishing robust evidence of genetic predisposition to lung cancer risk has a potential clinical utility for not only population risk stratification but also primary prevention. The main objective of our study was to identify, consolidate, and interpret genetic associations of common variants with lung cancer using a comprehensive research synopsis and systematic meta-analysis. We attempted to systematically evaluate all published candidate-gene association studies in lung cancer following credible guidelines, which were used to guide and standardize these field synopses. A total of 2,910 genetic variants from 754 unique candidate genes or loci were eligible for further analyses. The included studies had a mean of 414 cases (range 13–4257) and 565 controls (range 12–55823). Among the 2,910 variants, 254 were reported in at least three independent datasets, and eight had been reported as the top association variants with lung cancer (P < 5 × 10−8) in published GWAS24. Therefore, our meta-analyses were focused on the remaining 246 genetic variants in 138 genes or loci (Supplementary Table ). More detailed information of the variants was presented in the Supplementary Results.Among the final 1,018 eligible publications for our meta-analysis   and were combined from eight studies (range 3–133) (Supplementary Table 5) Table . The strAPEX1 rs1760944, AXIN2 rs2240308, CHRNA3 rs6495309, CXCR2 rs1126579, CYP2E1 rs6413432, HYKK rs931794, PON1 rs662, and REV3L rs462779) were strong and ten were moderate   Table .ATM rs66467 and REV3L rs465646) showed strong and moderate cumulative epidemiological evidence, respectively  showed moderate evidence . We found that 19 and 26 variants were significantly associated with lung cancer susceptibility in Caucasian and Asian population, respectively. Five variants  showed strong and five  showed moderate evidence in the Caucasian population  and seven variants , respectively  were unique to the Asian population, and seven  to Caucasian population. Four variants  shared between the two groups, including one (CLPTM1L rs402710) showed consistent strong evidence of significant associations in both groups  depending on the available data on Table . For thely Table . CompariCHRNA5 rs16969968, CLPTM1L rs402710, CYP2E1 rs6413432, ERCC1 rs11615, FGFR4 rs351855, HYKK rs931794, MIR146A rs2910164, and TERT rs2736098) demonstrated strong cumulative epidemiological evidence  shared between the NSCLC and SCLC group . As for SCC, two out of eight significant variants (CYP1A1 rs4646903 and CYP2E1 rs6413432) showed strong cumulative evidence. Four significant variants  were shared between the AD and SCC group, including one (CYP2E1 rs6413432) showed consistent strong evidence of significant associations in both groups   or weak cumulative evidence, mostly due to lack of large-scale evidence and the presence of potential biases (Table ERCC1 rs11615), six , and three variants, respectively. Comparing the significant variants between two groups, seventeen were unique to the smoking population, five to the non-smoking population, and five shared between the two populations , and the others in non-coding regions  , four , and 139 variants, respectively  in both main and subgroup meta-analyses  at the 5′-promoter region of APEX128 was shown as a significant variant  with strong cumulative evidence. This variant is predicted to influence promoter histone marks in 24 tissues including lung and lung cancer cell lines. Previous in vitro promoter assay has detected that the rs1760944 T allele significantly lowered promoter activity than that of the G allele, which indicated the variant allele (T) may be associated with a low transcriptional activity of the APEX1 in lung cancer cells28. The variant rs6495309 in CHRNA3/B4 intergenic region12 showed strong evidence of association with lung cancer susceptibility in our meta-analysis. This finding was consistent with the results from a previous meta-analysis performed in Chinese population29, and a recent meta-analysis performed on the basis of GWASs of lung cancer15. Additional subgroup analysis of Asians in our study also showed the risk effect for the rs6495309 C allele. This SNP overlaps with promoter histone marks and alters regulatory motif. Functional study also demonstrated that the rs6495309 C allele significantly increased the CHRNA3 expression through altering the ability of CHRNA3 promoter binding to the transcriptional factor Oct-112. A common genetic variation rs1126579 (C > T) located in the 3′UTR of the CXCR2 (IL8RB) was found to be associated with a reduced risk of lung cancer with strong evidence. The HaploReg tool identified that rs1126579 was an eQTL for a number of genes including CXCR2. Previous studies also reported that CXCR2 was down regulated in lung cancer tissue and might play a suppressive role in lung cancer via the p53-dependent senescence31. Functional data indicated that the rs1126579 variant can disrupt the binding site of miR-516a-3p and further increase the expression of CXCR230, which may also explain why rs1126579 showed a protective effect on the risk of lung cancer.Variants in non-coding region associated with lung cancer risk may have their effects through transcription, mRNA stability, protein structure/function, or binding sites of miRNA32. For example, the non-synonymous CHRNA5 rs16969968 (Asp398Asn) causes an amino acid substitution at codon 398 of the CHRNA5 protein. And the aspartic acid (Asp398) is located at the central part of the second intracellular loop in the structure of CHRNA5 protein, and was reported highly conserved across multiple species10. The rs1042522 (Arg72Pro) is a common functional SNP in the exon 4 of TP53, which encodes an important tumor suppressor protein. TP53 gene is often mutated in NSCLC tumors, an early event in development of lung cancer33. Further functional data showed that the 72Pro allele carriers of lung cancer patients may have a low frequency of the TP53 mutations in tumors34. The rs351855 (Gly388Arg) influences the transmembrane domain of the FGFR4 protein35. This SNP resides in a conserved region and causes a possibly damaging effect on protein function of FGFR4 predicted by PolyPhen. Also, rs4800  is a non-synonymous SNP in SOD2. This SNP with valine variation can reduce enzyme activity36 and further increase oxidative stress. Rs2736098 is a synonymous SNP (Asn305Asn) in exon 2 of the TERT gene, which is a well known oncogene and encodes the catalytic subunit of the telomerase37. This SNP may have association with telomere length38. Although it does not change protein amino acid, this SNP is located within the gene regulatory elements and may alter transcription factor binding.Variants falling within coding regions, especially non-synonymous SNPs, could have some effects on protein structure, function, or expression level, which may explain its association with the susceptibility of diseaseMIR146A and rs11614913 (C > T) in the mature sequence of miR-196a-3p encoded by MIR196A239. Both SNPs showed significant miRNA expression differences between their alleles40 and could affect the stability of secondary hairpin structure39. Study also showed that rs2910164 can influence the interaction between miR-146a-3p and its potential target genes, and rs11614913 can increase the affinity of miR-196a-3p for TP5339.In addition, we found two SNPs with strong evidence of associations with lung cancer risk are located in miRNA gene coding regions, rs2910164 (C > G) in the seed of miR-146a-3p encoded by 41 and indicated that some variants  with strong evidence may be ethnic-specifically associated with lung cancer risk. Previous studies had demonstrated the existence of different genetic background in different histological subtypes of lung cancer42. When cases were stratified according to histological types, the associations between several variants  and specific subtypes of lung cancer were of strong evidence. A growing number of studies demonstrates interactions between genetic variants and smoking44. Our subgroup analysis also found that some variants showed significant associations with lung cancer risk in smokers but not in non-smokers, for example CYP1A1 rs4646903 and GSTP1 rs1695.Our subgroup analyses also provided additional important details of genetic associations in specific groups. The results of subgroup meta-analyses by ethnicity supported the well-known cognition of “racial” differences in genetic effects for complex diseases including lung cancerAs the purpose of meta-analysis is not only to reveal genetic variants significantly associated with lung cancer risk, but also to identify the variants with non-significant associations. Our study revealed that 150 variants in 98 genes had non-significant associations with lung cancer risk. However, most of these variants had weak cumulative epidemiological evidence due to the presence of insufficient statistical power (119/150) and/or strong between-study heterogeneity (73/150), and only 11(7.3%) variants had strong or moderate cumulative evidence. Our results provided important clues to further assess the main effects of these variants.I2 > 50) was more than 40% for all meta-analyses with a significant result. Although subgroup analyses stratified by ethnicity, histology, and smoking status were performed to address the heterogeneity, other sources of heterogeneity could exist and are difficult to address because of limited available data. Third, although we tried to explore the consistency and difference in genetic associations between some variants and lung cancer risk across different ethnic groups, meta-analyses stratified by ethnicity were performed only for Caucasian and Asian populations. Since very few enrolled original studies were carried out in other descent populations (e.g. African descent), the available data were not sufficient to perform subgroup meta-analyses in other descent populations. Additional association studies are needed to establish in populations of other ethnic descent for these reported variants. Finally, although we conducted systematic evaluations of cumulative epidemiological evidence for variants associated with lung cancer risk, biases cannot be completely excluded in this study.Despite a comprehensive and systematic approach was applied to the synopsis of genetic association studies in lung cancer, several limitations should be considered when interpreting our results. First, although available studies were searched widely and eligible studies were selected strictly according to the inclusion and exclusion criteria, it is possible that some studies might have been overlooked. Our studies didn’t include research published in the form of abstracts or in language other than English. However, for most abstracts, we also searched and included relevant studies published with whole text and reported by the same research groups. Publication biases were not identified in most meta-analyses with significant association results. Also, the proportion of studies published in language other than English is small therefore it should not have significant influence on the main results. Second, the percentage of meta-analyses with high heterogeneity  guiding genetic reviews specifically, and Meta-analysis Of Observational Studies in Epidemiology (MOOSE) guidelines−8) identified by GWAS since they have been replicated by many studies.A study for inclusion had to meet the following four criteria: (1) it evaluated the association between a genetic polymorphism and lung cancer risk using a case-control, cohort, or a cross-sectional design in human; (2) lung cancer cases were diagnosed by pathological and/or histological examination; (3) it was published in a peer-reviewed scientific journal or online in English; (4) it provided sufficient information of genotype and/or allelic distributions for both cases and controls. We excluded studies with a family-based design and loci with genome-wide significant  were also separately extracted from each study whenever possible. Ancestry was divided into four general groups  based on ancestry of at least 80% of the subjects41. If no details of ethnicity were reported, the determination was made based on the general population of the country or region where the study was done41. When a publication reported data from multi-racial groups, data for each population were extracted and analyzed separately if possible.When multiple publications used the same or overlapping data sets, we kept the data with the largest population or most recent ones as recommended by Little http://www.ncbi.nlm.nih.gov/projects/SNP/index.html), to designate the reported variants. For articles with “rs” number, we used as it was; for these without we used bioinformatics tools such as NCBI Blast (http://www.ncbi.nlm.nih.gov/BLAST/) and UCSC In-Silico PCR (http://genome.ucsc.edu/cgi-bin/hgPcr) to find “rs” number for the reported variant; for the remaining without any “rs” number, we used the common nomenclature  in the original articles.To avoid the variant nomenclature confusion from different articles, we used the most current gene names and uniform identifiers (“rs” number) of variants in a public single nucleotide polymorphism (SNP) database , except where indicated otherwise. All tests were two-sided and considered statistically significant when p value was at 0.05 or lower, unless otherwise stated.18. Association between a variant and lung cancer risk was assessed by study-specific crude odds ratios (ORs) and 95% confidence intervals (CIs) using a DerSimonian and Laird random-effects model47. The initial main meta-analyses assessed the variant effect using an allelic genetic model  without stratification. For the variation not in the form of single nucleotide substitution, a conventional comparison from the publications was used to assess the effects . When average minor allele frequency (MAF) were greater than 50%, a rare occasion where major and minor alleles are flipped in different ethnic populations, we designated the minor allele from Caucasian population in all analyses. For the variant with sufficient genotype distribution data, we performed additional analyses based on dominant and recessive genetic models.All variants from at least three data sources were selected for meta-analysisSubgroup meta-analyses were also performed by ethnicity (Caucasian and Asian), histological types , and smoking status (smoking and nonsmoking), if sufficient data were available.48. We also used I2 heterogeneity metric to assess the heterogeneity49. Generally, I2 < 25%, 25%-50% and > 50% showed mild, moderate, and strong heterogeneity, respectively.Between-study heterogeneity was assessed by calculating the Cochran Q statistic, with a p value less than 0.10 being the significant threshold50. Potential small study effect  was checked by the modified Egger’s test, which can lower the type I and type II error rates compared to the original Egger’s test51. We also conducted an excess significance test to examine whether there was a relative excess of formally significant findings in studies due to potential sources of bias, such as selective analyses, selective outcome reporting, or fabricated data52.The publication bias of studies was evaluated by funnel plot analysis (logOR against standard error) and Begg’s testFor all variants that showed a significant association with lung cancer risk, we performed a sensitivity analysis to examine whether the significant summary ORs were robust after excluding the first published or first positive report, or excluding studies with controls violating Hardy-Weinberg equilibrium [HWE]. We used a Fisher’s exact/chi-square to assess the HWE among controls in each dataset.21. Venice criteria is a semi-quantitative index which assigns three aspects for the amount of evidence, extent of replication, and protection from bias, and finally generates a composite assessment of “strong”, “moderate”, or “weak” epidemiological credibility for an association with lung cancer risk21. For the three aspects  of Venice criteria, each aspect was assigned three levels 21. Briefly, amount of evidence, depending on total sample size of the smallest genetic group among cases and controls in each meta-analysis, was graded as A (sample size >1000), B (sample size between 100 and 1000), or C (sample size <100). For very rare variant with frequency less than 0.5%, the amount of evidence was not assessed considering an A grade was unlikely to obtain18. The extent of replication, depending on between-study heterogeneity, was graded as A (I2 < 25%), B (I2 between 25% and 50%), or C (I2 > 50%). The protection from bias, considering various potential sources of bias in meta-analysis, was graded as A when there was no demonstrable bias and the bias would unlikely invalidate the association, B when there was insufficient information for identifying evidence  although there was no obvious bias, and C when the bias was evident and/or was likely to explain the presence of association. More specifically, C grade was assigned if the meta-analysis had any of the following potential sources of bias: (1) the magnitude of the association was low  with the exception of a highly consistent OR across studies enrolled in meta-analysis; (2) the sensitivity analysis indicated that the significant summary OR can be substantially changed; (3) the potential small study effect was present according to the modified Egger’s test ; (4) an excess of significant findings was possible ; (5) there was a potential publication bias . With the grades from three aspects, the credibility of cumulative epidemiological evidence was categorized as strong , moderate , or weak (any grade was C).For each nominally significant results from the meta-analyses, Venice criteria was used to assess the credibility of cumulative epidemiological evidence53. The credibility of cumulative epidemiological evidence of non-significant association was categorized as strong , weak , or moderate (for other cases).Additionally, for the non-significant associations revealed by all meta-analyses, we also evaluated the credibility of cumulative epidemiological evidence based on three aspects: the degree of heterogeneity across studies, potential bias assessment, and statistical power. The statistical power was calculated by using SNP toolsAll data generated or analysed during this study are included in this article and its Supplementary Information file.Supplementary information file


PPARG) plays an important role in the pathogenesis and maintenance of essential hypertension (EH). It has been suggested that polymorphisms of PPARG are associated with the risk of EH. However, findings to date remain controversial. To elucidate the associations between the PPARG Pro12Ala and C161T polymorphisms and EH risk, a meta-analysis was carried out.Peroxisome proliferator-activated receptor gamma , VIP and Wanfang databases was conducted. The pooled odds ratios (ORs) and 95% confidence interval (CI) were calculated to estimate the size of the effect using the random-effects model. At the same time, the pooled standardized mean difference (SMD) with 95% CI was used for the meta-analysis of the PPARGPro12Ala polymorphism and EH risk, were included in this study. Overall, the results suggested that Ala allele was associated with the decreased EH risk . The subgroup analysis stratified by ethnicity showed that the significant association between the PPARG Pro12Ala polymorphism and EH was only detected in the Asian subgroup. There was no difference in blood pressure values between Ala carriers and non-carriers. For the C161T polymorphism, only 5 studies comprising 1,118 cases and 1,357 controls met the inclusion criteria. The overall results showed that the PPARG C161T polymorphism was not associated with the risk of EH. But in the subgroup analysis, we found that the PPARG C161T polymorphism significantly associated with the risk of EH in the Asian subgroup .Finally, Fifteen papers (seventeen studies) including 4,151 cases and 4,997 controls to evaluate the association of the PPARG polymorphisms might be associated with the risk of EH.Our meta-analysis suggested that the  Essential hypertension (EH) is one of the leading causes of mortality and morbidity. Epidemiological data show that there are about one billion EH patients in the world. Pathogenesis of EH is related to multiple risk factors, including environmental and genetic factors , 2. SystPPARG, also known as NR1C3), a member of the nuclear hormone receptor subfamily, regulates the expression of a network of genes involved in adipogenesis and lipogenesis, insulin sensitivity, inflammation and atherosclerosis [PPARGs consist of three different isoforms-PPARG1, PPARG2 and PPARG3-result different promoter and different splicing methods. PPARG is mainly expressed in adipocyte tissue, as well in vascular endothelial cells, smooth muscle cells and monocyte/macrophage cells [PPARG plays an important role in adipose differentiation and in susceptibility to type 2 diabetes [PPARG levels have been reported in subjects with high blood pressure not only in vitro but also in vivo [Peroxisome proliferator-activated receptor gamma . The cytosine is substituted for guanosine at nucleotide position 34 (C34G), leading to a change from proline (Pro) to alanine (Ala) at position 12 of exon 2 in PPARG. This substitution may lead to the change of PPARG protein structure, which can decrease the effect of binding to the target gene and reduce its transcriptional activity [To date, numerous studies have been conducted to explore the relationship between activity , 14, whiPPARG Pro12Ala polymorphism with EH risk have been extensively performed previously, but the results remain controversial [vs. 6.92%, P = 0.001). Several other studies reached similar conclusions [PPARGPro12Ala polymorphism and EH. More contrary results were found in other studies [Studies on the association of oversial , 16, 17.oversial  exploredclusions , 19. Howclusions  did not  studies –24. In 2 studies  examined studies  concluders3856806), which is located in the exon6 of PPARG. As for association between the C161T polymorphism and the risk of EH, the results are also inconsistent. For example, Qu et al [vs. 36.2%, P<0.05) and that the T allele protected one from EH. In 2014, Zhang et al [PPARG C161T polymorphism was not associated with EH susceptibility [Another variant is C161T  and (“polymorphism” or “variant” or “gene” or “mutation”) and . For example, the full search strategy used in the PubMed database is: text word = (“PPAR” or “peroxisome proliferator-activated receptor” or “NR1C3”) and (“polymorphism” or “variant” or “gene” or “mutation”) and . To find additional eligible studies, the reference lists of the studies included were searched manually.The meta-analysis followed the Preferred Reporting Items for Systematic Reviews and Meta-analysis criteria . A comprPPARG polymorphisms and EH risk; 2) Case-control study; 3) Having clear, original data of genotypic and/or allelic frequencies; 4) Studies written in English and Chinese; 5) Hypertension being defined as systolic blood pressure ≥140 mmHg and/or diastolic blood pressure ≥90 mmHg or treatment with anti-hypertensive medication; 6) For blood pressure, the data being presented as the mean± standard deviations (SD).Eligible studies had to meet the following inclusion criteria: 1) Studies on the association between The exclusion criteria were as follows: 1) a review or case report; 2) genotypic and allelic frequencies were not clear; 3) animal studies; 4) secondary hypertension; 5) cross-sectional or cohort study; 6) data published repeatedly.A special table was used to record the following information: the name of first author, year of publication, country, ethnicity, age, type of study, source of control, genotyping methods, the distribution of genotypes in cases and controls, diagnostic criteria of EH, and the value of blood pressures. The original data were extracted by two investigators (Cai GJ and Zhang BF) independently. Disagreements were resolved by consulting with a third author (Weng WJ). If we were in doubt about the data in the study, we tried our best to contact the author for correspondence by email.PPARG genotype distributions of control groups was checked by Fisher’s exact test. Pooled odds ratios (ORs) and 95% confidence interval (CI) were used to assess the strength of association between PPARG polymorphisms and EH risk, and the pooled standardized mean difference (SMD) with 95% CI was used for the meta-analysis of PPARG polymorphisms and blood pressure. The between-study heterogeneity was evaluated by the Chi-square-based Cochrane’s Q statistic and the I2 statistic. If the between-study heterogeneity was significant , a random effect model (Dersimonian-Laird method) was used to calculate the result; otherwise, a fixed effect model (Mantel-Haenszel method) was used [vs. CC; for C161T, TT+TC vs. CC) and allelic  model were performed to calculate the pooled ORs in the present meta-analysis. Subgroup analyses were also conducted by ethnicity (Caucasian and Asian), according to HWE (Yes and No), source of control  and genotyping methods (PCR-RFLP and others) to explore the source of heterogeneity. A sensitivity analysis omitting an individual study each time was used to evaluate the stability of the main meta-analysis results. The potential publication bias was examined by Begg’s funnel plot, and the funnel plot asymmetry was assessed by Egger’s linear regression test. AP-value <0.05 (two-sided) was considered statistically significant.We used Stata 12.0  software for all the statistical analysis. Hardy-Weinberg equilibrium (HWE) for the was used . Due to Fig 1 lists the flow diagram of the article selection process. The 17 studies (15 papers) evaluated the PPARG Pro12Ala polymorphism and EH risk, including 4,151 cases and 4,997 controls [PPARG C161T polymorphism and EH risk, including 1,118 cases and 1,357 controls [Tables 2 summarise the characteristics of included articles, including ethnicity, year of publication, genotyping method, distribution of genotypes and alleles, etc. The diagnostic criteria of EH were appropriated in all of these studies. The controls in 3 studies deviated from HWE [PPARG Pro12Ala polymorphism and the value of systolic blood pressure (SBP) and diastolic blood pressure (DBP) (Table 3). Of those, five papers were conducted in Asians and one paper in Caucasians. Because only two papers studied the relationship between the PPARGC161T polymorphism and blood pressure values, we did not perform meta-analysis.A total of 412 potentially relevant papers were obtained by the literature search, of which 17 papers met the inclusion criteria. controls , 29–34. controls , 33 contcontrols –26, 34. from HWE , 32. SixTable 4 summarises the association between the PPARGPro12Ala polymorphism and the risk of EH. Because between-study heterogeneity was significant in the overall analysis, a random-effect model was used. There were significant associations between the PPARG Pro12Ala polymorphism and EH risk in the whole population under allelic  and dominant genetic models  .PPARG Pro12Ala polymorphism and EH was only detected in Asian population under allelic  and dominant genetic models  .In the subgroup analysis stratified by ethnicity, the significant association between PPARG Pro12Ala polymorphism was associated with EH in a population-based subgroup  and dominant genetic models  .In the subgroup, based on whether the control genotype distribution displayed HWE, a significant association was found between the PPARG Pro12Ala polymorphism and EH risk in the PCR-RFLP subgroup, whereas a statistically significant association was found in the other methods subgroup  .Stratification based on the genotyping method revealed no significant association between the S3 Fig, there was no significant difference in the value of blood pressure between Ala carriers and non-carriers .As shown in Fig 5, there was no significant association between the PPARG C161T polymorphism and EH risk in the whole population under allelic  and dominant genetic models .As shown in I2 = 75.0%, P = 0.003; for dominant model, I2 = 78.9%, P = 0.001), so subgroup analyses stratified by ethnicity, genotyping methods and source of control were also conducted. The subgroup analyses showed that the statistically significant association only existed in the Asian subgroup  .Because the between-study heterogeneity was also significant in the whole study . When the Grygiel-Górniak et al. study was excluded, the pooled result changed to the contrary.Because of the significant between-study heterogeneity, sensitivity analysis was conducted to determine the influence of individual studies on pooled ORs by sequentially removing each eligible study. No single study of the Pro12Ala polymorphism influenced the stability of the entire study . For theFig 9 showed that a visual publication bias was found in the Begg’s funnel plot under both allelic and dominant models. The significant difference was also detected by Egger’s test . For the C161T polymorphism, no visual publication bias was found in the Begg’s funnel plot under both of the genetic models, which was confirmed by the Egger’s test . No obvious publication bias was found in the Begg’s funnel plot for blood pressure , which was also confirmed by Egger’s test .The Begg’s funnel plot and Egger’s test were adopted to evaluate the publication bias among the individual studies. For Pro12Ala, PPARG polymorphisms and the risk of EH. Our meta-analysis found associations of the Pro12Ala and C161T polymorphisms of PPARG with significant risk of EH.In the present meta-analysis, we evaluated the relationship between PPARG modulates the expression of genes involved in energy storage and utilisation [Studies suggest that lisation  and playPPARG might be associated with EH risk and the activation of central°PPAR-γ could attenuate angiotensin II-induced°hypertension [PPARG decreases blood pressure is still not completely clarified. The likely mechanisms are: 1) antagonism of the renin-angiotensin system, 2) inhibition vascular cell proliferation; and 3) improvement endothelial function [Clinical and experimental studies have indicated that rtension . The mecfunction .PPARG had been studied, including Pro12Ala (rs1805192), C161T (rs3856806), C681G (rs10865710). As one of the important candidate genes for EH, studies on the association of PPARG polymorphisms and EH risk have been explored extensively. The results showed that PPARG polymorphisms might be associated with metabolism-related diseases [PPARG when it was in accord with the included criteria and was studied more than 3 times. After a literature search, the Pro12Ala and C161T polymorphisms in PPARG matched these conditions.Up to now, several polymorphisms in  C681G rs865710. APPARGPro12Ala and C161T polymorphisms and EH have been conducted so far, the results of individual studies were inconsistent. In 2003, a community-based, cross-sectional observational study to explore the relationship between the PPARG Pro12Ala polymorphism and blood pressure in male subjects with type 2 diabetes was performed in Swedish subjects. Subjects with the Pro/Ala or Ala/Ala genotype had significantly lower diastolic blood pressure. On the contrary, in a cross-sectional study conducted in the aboriginal Qatari population in 2013, Beneret al. [Although much research on the relationship between eret al.  showed teret al.  and Uygheret al. , rather eret al.  and Zhuaeret al. .PPARG Pro12Ala polymorphism was associated with susceptibility to EH under both allelic and dominant models, which was consistent with the study by Wang et al [PPARGPro12Ala polymorphism. Subgroup analysis stratified by ethnicity indicated that the association existed in Asians, but not Caucasians. However, there was no significant difference in the value of blood pressure between Ala carriers and the non-carriers. There was no significant association between the PPARG C161T polymorphism and EH risk in the whole population.However, in the subgroup analysis, we found that the PPARGC161T polymorphism was significantly associated with the risk of EH in the Asian subgroup. Sensitivity analysis showed that the Grygiel-Górniak et al. study significantly affected the overall results. When this study was excluded, the pooled result changed to the contrary. When we analysed this paper carefully, we found that this study was conducted in Caucasians and that the sample size was the smallest among the 5 papers, which contributed to the false-negative result.In the present study, the results demonstrated that the ng et al . Althougrs1801282) in PPARG and EH susceptibility. They concluded that the Ala allele might be protective for hypertension among East Asians, but not among Caucasians. Compared with the study by Wang et al, the present meta-analysis had several strengths. First, not only Pro12Ala, but also C161T polymorphisms in PPARG were involved in this study, and the analysis of correlations between polymorphisms and EH were more comprehensively studied. Second, only 8 papers with 3,281 participants were involved in the study by Wang et al. in 2012. Since then, several new case-control studies on the relationship between PPARGPro12Ala polymorphism and the susceptibility to EH were conducted. In the present meta-analysis, a total of 15 articles comprising 4,151 cases and 4,997 controls were involved, which increased the statistical power greatly. Third, due to the significant between-study heterogeneity, we performed sensitivity analysis and subgroup analysis to explore the source of heterogeneity.In 2012, Wang et al.  conductePPARG C161T polymorphism and susceptibility to EH. The sample size was relatively small, and any conclusion should be made cautiously. Finally, the publication bias was significant for the Pro12Ala polymorphism. A likely reason for the publication bias was that studies in other languages were not retrieved, and/or the negative studies were less likely to be published.Several limitations are present in our meta-analysis. First, as mentioned above, the association between Pro12Ala polymorphism and EH might be gender-specific. Because we could not obtain individual data from each study, we could not analyse the interaction between gene polymorphisms and the environment. Second, in our meta-analysis, most of the studies were conducted in Asians. There were only three studies conducted in Caucasians. Since the sample size for Caucasians was particularly small, the power was low, and thus results should be interpreted with caution. Third, only five studies with a small sample size  were involved in our analysis of the association between the PPARGPro12Ala and C161T polymorphisms might be associated with the risk of EH in Asians. Since the Caucasian sample size was too small and statistically lacking in power, we should interpret it cautiously. Further large-scale studies should be conducted to confirm the above conclusions.In conclusion, our meta-analysis suggested that the S1 File(DOC)Click here for additional data file.S2 File(DOCX)Click here for additional data file.S1 FigPPARG Pro12Ala polymorphism and susceptibility to EH in subgroup analysis based on HWE under the allelic model (A) and the dominant model (B).Forest plots of ORs for the association between the (TIF)Click here for additional data file.S2 FigPPARG Pro12Ala polymorphism and susceptibility to EH in subgroup analysis based on genotyping methods under the allelic model (A) and the dominant model (B).Forest plots of ORs for the association between the (TIF)Click here for additional data file.S3 FigPPARG Pro12Ala polymorphism and the value of systolic blood pressure (A) and diastolic blood pressure (B).Forest plot of the association between the (TIF)Click here for additional data file.S4 FigPPARG C161T polymorphism and the value of systolic blood pressure (A) and diastolic blood pressure (B).Funnel plot for the (TIF)Click here for additional data file.


P < 0.001] and childhood obesity . Antibiotic exposure in early life also significantly increased the z-score of childhood body mass index . Importantly, there was an obvious dose–response relationship between antibiotic exposure in early life and childhood adiposity, with a 7% increment in the risk of overweight  and a 6% increment in the risk of obesity  for each additional course of antibiotic exposure. In conclusion, antibiotic exposure in early life significantly increases risk of childhood obesity. Moreover, current analyses are mainly taken from developed countries, and therefore the impact of antibiotic exposure on risk of childhood obesity in vulnerable populations or developing countries still needs to be evaluated in future studies.A number of studies have previously assessed the impact of antibiotic exposure in early life on the risk of childhood obesity, but no systematic assessment is currently available. A systematic review and meta-analysis was performed to comprehensively and quantitatively elucidate the risk of childhood obesity caused by antibiotic exposure in early life. Literature search was performed in PubMed, Embase, and Web of Science. Random-effect meta-analysis was used to pool the statistical estimates. Fifteen cohort studies involving 445,880 participants were finally included, and all those studies were performed in developed countries. Antibiotic exposure in early life significantly increased risk of childhood overweight [relative risk (RR) = 1.23, 95% confidence interval (CI) 1.13–1.35,  Obesity has increased significantly among children and adolescents over the past several decades, and childhood adiposity has become a major challenge to public health worldwide , 2. In tEmerging evidence has suggested that gut microbiota is linked to obesity and other metabolic disorders , 12. TheA literature search in PubMed, Emabse, and Web of Science was carried out to find articles that examined the impact of antibiotic exposure on childhood adiposity risk. The following terms were used:  AND (adiposity OR obesity OR obese OR overweight OR body mass index) AND (child OR children OR childhood OR boy* OR girl* OR infancy OR infants). All databases were searched from their inception through September 6, 2016, and an updated literature search was performed on February 26, 2017. Language restriction was not applied. Bibliographies of eligible articles and relevant reviews were also screened.z-scores of childhood BMI or weight between the exposed group and the non-exposed group. Infancy is usually defined as the period from birth to 2 years of age. Early life includes infancy and prenatal period, and is thus from fetal life through 2 years of age in the present study. Studies containing overlapping data, case reports, or case series were all excluded.Selection criteria for the meta-analysis were as following: (1) prospective or retrospective cohort studies; (2) the outcomes of interest were the association between antibiotic exposure and childhood adiposity, or the impact of antibiotic exposure on childhood body mass index (BMI) or weight; (3) the exposure time was early life including infancy and prenatal period; (4) reporting risk estimates on the associations, such as relative risks (RRs) with 95% confidence intervals (95% CIs), or the difference in the Two colleagues extracted data independently, and discrepancy was resolved by discussion. The extracted data mainly included first author, country, study design, antibiotic exposure, participants, follow-up time, adjusted variables, and outcomes of interest. If one study reported data of multiple follow-ups, we used the data for the longest follow-up time. The authors of included studies were contacted if crucial information was not provided in the articles.z-scores of childhood BMI or weight between the exposed group and the non-exposed group. The evaluation of study quality was performed by the Newcastle–Ottawa scale, and studies scoring seven or more points were identified as high-quality studies . Meta-analysis showed that antibiotic exposure in early life  significantly increased the risk of childhood overweight  . Meta-analysis showed that antibiotic exposure in early life  also independently increased childhood obesity risk  . Subgroup analysis by exposure time suggested that antibiotic exposure during either infancy or prenatal period both increased childhood obesity risk (Table There was moderate heterogeneity among studies relating antibiotic exposure and childhood obesity I = 46.8%.) Figure , antibioz-score (I2 = 7%). Meta-analysis showed that antibiotic exposure in early life could dramatically increase the z-score of childhood BMI  .Meta-analysis of two studies , 22 relaPEgger’s test was 0.99 and it also proved the lack of publication bias. In addition, funnel plots in other meta-analyses on childhood obesity or childhood BMI also did not show an obvious risk of publication bias.The funnel plot did not show an obvious risk of publication bias Figure . PEgger’P = 0.03) and a 6% increment in the risk of childhood obesity  for each additional course of antibiotic exposure (Figures Dose–response meta-analysis of six studies , 41, 45 z-scores of childhood BMI and body weight. Thus, the meta-analysis provides strong evidence for the impact of antibiotic exposure during early life on childhood adiposity risk.The present study is the first systematic review comprehensively evaluating the impact of antibiotic exposure on childhood adiposity. Fifteen cohort studies involving 445,880 participants were finally included , 49. Thez-scores of childhood BMI or body weight from included studies were also adjusted for confounding factors, and the pooled z-scores were statistically significant (Table Another important issue is whether antibiotic exposure is an independent risk factor for childhood adiposity. In present meta-analysis, all included studies provided risk estimates adjusted for confounding factors Table . The poont Table . Therefoz-score in children with overweight or obesity (The pathogenesis of obesity is complex and has not been well understood, and it is the same with childhood obesity , 51. The obesity .Data from animal studies have shown that antibiotic-induced changes in gut microbiota can result in fat accumulation by changing host metabolism . AntibioThe increasing prevalence of childhood adiposity has become a major global health challenge, and actions to reduce the prevalence of obesity in children and adolescents are urgently needed , 63, 64.Apart from the prudent use of antibiotics in infants, it is also important for us to develop some plausible methods to prevent obesity in those infants or children who have to receive the treatment of antibiotics. Several studies have studied some possible ways to prevent obesity in children who expose to antibiotics. Kaliannan et al. demonstrated a beneficial effect of omega-3 fatty acid in preventing antibiotic-induced gut dysbiosis and obesity . EconomoThe meta-analysis had several strengths. Firstly, the large number of participants could provide a precise evaluation of the relationship between antibiotics and childhood adiposity. In addition, the obvious dose-dependent relationship strengthened the evidence for the association. Finally, the consistency in the findings of several different types of outcomes further supported the robustness of the findings.There were several limitations in the meta-analysis. Firstly, all of the current studies were taken from developed countries and there were missing data from developing countries or vulnerable populations. The current study is representative of developed countries, whereas the impact of antibiotics in children in vulnerable populations or developing countries may likely be different from that found in developed countries. Therefore, the impact between antibiotic exposure and childhood adiposity in developing countries or vulnerable populations is still unclear, and needs to be evaluated in future research. Secondly, there were obvious differences in those included studies. The confounding factors used in the adjusted estimates were different, which could result in heterogeneity and different outcomes . ThirdlyIn summary, this meta-analysis provides strong evidence for the association between antibiotic exposure in early life and childhood adiposity. There is also an obvious dose–response relationship between antibiotic exposure and childhood adiposity. Prudent use of antibiotics is recommended for infants and children to reduce the risk of childhood adiposity. Future studies are also required to develop some effective methods to prevent obesity in infants or children exposing to antibiotics.XS and JZ designed the study. XS, BW, XD, RS, and JZ contributed to the literature search, interpretation, writing, and proofreading of the manuscript. XS, BW, LL, and QY extracted data and performed data analyses. XS, BW, and XA revised the study. XS and BW generated the figures.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


I2 = 33.6%). The subgroup analysis indicated that the protective function of dietary calcium intake were observed in esophageal squamous cell cancer, but not in esophageal adenocarcinoma in the studies conducted in Asia, but not those in Europe and America. In conclusion, our results suggest that higher dietary calcium intake is associated with a lower risk of esophageal cancer—especially esophageal squamous cell cancer—in Asian populations, though more data from prospective cohort studies are needed.Although several epidemiological studies have investigated the association between dietary calcium intake and the risk of esophageal cancer, the results are inconsistent. This study aimed to make a comprehensive evaluation regarding the association between calcium intake and risk of esophageal cancer through a meta-analysis approach. We searched for all relevant articles from the inception to April 2017, using PUBMED, EMBASE, and Web of Knowledge. The pooled odds ratio (ORs) with the 95% confidence interval (95% CI) for the highest versus the lowest categories of calcium intake was calculated using a Mantel–Haenszel fixed-effect model. In total, 15 articles reporting 17 studies including 3396 esophageal cancer cases and 346,815 controls were selected for the meta-analysis. By comparing the highest vs. the lowest levels of dietary calcium intake, we found that dietary calcium intake was inversely associated with the risk of esophageal cancer (OR = 0.80, 95% CI: 0.71–0.91,  Esophageal cancer (EC) is one of the most common cancers in the world. With estimated 455,800 new cases and 400,200 deaths in 2012, esophageal cancer has been the tenth most common malignancy and the eighth leading cause of cancer-related deaths [Epidemiologic evidence demonstrated a number of risk factors for esophageal cancer, such as age, gender, alcohol drinking, tobacco smoking, obesity, chronic gastroesophageal reflux disease, dietary carcinogens, and insufficiencies of micronutrient consumption ,4,5, amoThree electronic databases  were searched aiming to assess the association between dietary calcium intake and esophageal cancer up to April 2017. The following search terms were used:  AND  AND (cancer OR tumor OR tumour OR carcinoma OR neoplasm). Besides, the references cited within the related articles were also searched for additional eligible publications. Only full-text original journal articles with a cohort or case–control study design were included.To be included in this meta-analysis, the studies had to meet the following inclusion criteria: (1) the study design was a cohort or case–control study; (2) the exposure of interest was dietary calcium intake; (3) the outcome was esophageal cancer; (4) relative risk (RR), hazard ratio (HR), or odds ratio (OR), and corresponding 95% confidence intervals (95% CI) for the highest versus the lowest calcium intake were reported or could be calculated. If data were duplicated in more than one study, the one with the largest number of cases or the longest follow-up period was selected.Two reviewers independently assessed the articles with the inclusion criteria and extracted data with a standardized form , and anyI2 statistics. Generally, for the Q-test, heterogeneity with a value of p < 0.05 was considered as statistically significant. For the I2 statistic, the following cut-off points were used: <25% (low heterogeneity), 25–50% (moderate heterogeneity), and >75% (severe heterogeneity). If p < 0.05 and I2 > 50%, a DerSimonian and Laird random-effect model was used. Otherwise, a Mantel–Haenszel fixed-effect model was applied [The pooled ORs with corresponding 95% CI  were calculated to measure the association across studies. The heterogeneity among studies was examined by Q-test and  applied . p < 0.10) [p-values were two-tailed, and p < 0.05 was considered as significant, unless explicitly stated.Meta-regression and subgroup analyses were performed to explore the possible source of heterogeneity based on geographic location , study design (population-based case–control (PBCC), hospital-based case-control (HBCC), and cohort), pathological type , dietary assessment , publication year (before/in 2000 or after 2000), and adjustment for energy intake/body mass index (yes or no) . The “le < 0.10) . The staThe study selection process and the results are shown in I2 = 33.6%), suggesting that calcium intake was inversely associated with the risk of esophageal cancer.As shown in p > 0.05). Next, we conducted meta-regression analysis and subgroup analyses by geographic location, study design, pathological type, dietary assessment, publication year, as well as adjustment for energy intake and BMI. Meta-regression analysis showed that the inverse association between calcium intake and esophageal cancer was not significantly affected by these factors (I2 = 0.0%), published after 2000 , and adjusted for dietary energy intake , with no statistically significant heterogeneity. Likewise, negative association was also found in studies of ESCC , cohort studies , and studies with high quality score , as well as in studies using validated method . The heterogeneity alleviated in all four subgroups. The inverse association was not altered in either subgroup when stratified by adjustment for BMI. Heterogeneity decreased in the subgroup adjusted for BMI, while it increased in the other subgroup.As shown in I2 = 0.0%, p = 0.489). Egger’s test showed no evidence of significant publication bias in this meta-analysis . The funnel plot was provided in Sensitivity analysis was conducted by leaving one study out in turn and pooling the ORs of the remaining studies. The summary ORs did not substantially change, which indicated that our results were statistically robust. However, after excluding the study conducted by Graham , there wThis study is the first systemic meta-analysis regarding the association between dietary calcium intake and the risk of esophageal cancer, which is based on 15 articles reporting 17 studies (3 cohort studies and 14 case–control studies), including 3396 cases and 346,815 controls. The sample size is large enough to evaluate the effect of calcium intake on esophageal cancer.In addition to its bone formation role, calcium is also a ubiquitous second messenger and plays a critical role in human health . It was However, some information should be considered in the subgroup analyses. First and foremost, in the subgroup analysis by geographic locations, we found the inverse association between dietary calcium intake and esophageal cancer in studies of Asia, but not in studies of America and Europe, which indicated that the results are acceptable to Asian populations, but cannot be extended to Americans and Europeans. According to the information about calcium intake quantity presented in The present meta-analysis has several advantages. First, our analysis is the first comprehensive meta-analysis to reveal the possible associations between dietary calcium intake and risk of esophageal cancer. Second, it enables a reliable conclusion by using meta-regression analyses and sub-group analyses to explore the sources of heterogeneity. However, several limitations should also be noted. First, most of the eligible studies were case–control studies, which were difficult to rule out the influence of recall bias. Second, the levels of dietary calcium intake ranged widely among the studies included in our meta-analysis, and the definitions of the intake categories were different. Another limitation is that this meta-analysis is study-based, but not individual patient-based. Finally, we did not perform a dose–response effect analysis due to the incomplete data of dietary calcium intake. The present study suggests that a higher intake of dietary calcium might have protective effect against esophageal cancer—especially esophageal squamous cell cancer—in Asian populations. To further solidify the association of dietary calcium intake with the risk of esophageal cancer, well-designed studies—especially prospective cohort studies with validated FFQ and adjusted for dietary energy intake—should be conducted.


ERCC2) gene plays important roles in the nucleotide excision repair pathway. There is contrasting evidence on the association between the ERCC2 Asp312Asn polymorphism and the risk of cancer. We conducted a comprehensive meta-analysis in order to assess the correlation between these factors. We searched the PubMed, EMBASE, Science Direct, Web of Science, and CNKI databases for studies published from January 1, 2005 to January 1, 2016. Finally, 86 articles with 38,848 cases and 48,928 controls were included in the analysis. The overall analysis suggested a significant association between the ERCC2 Asp312Asn polymorphism and cancer risk. Furthermore, control source, ethnicity, genotyping method, and cancer type were used for subgroup analysis. The result of a trial sequential analysis indicated that the cumulative evidence is adequate; hence, further trials were unnecessary in the overall analysis for homozygote comparison. In summary, our results suggested that ERCC2 Asp312Asn polymorphism is associated with increased cancer risk. A significantly increased cancer risk was observed in Asian populations, but not in Caucasian populations. Furthermore, the ERCC2 Asp312Asn polymorphism is associated with bladder, esophageal, and gastric cancers, but not with breast, head and neck, lung, prostate, and skin cancers, and non-Hodgkin lymphoma. Further multi-center, well-designed studies are required to validate our results.Cancer is the leading cause of death in economically developed countries and the second leading cause of death in developing countries. The relationship between genetic polymorphisms and the risk of cancers has been widely researched. Excision repair cross-complementing group 2 ( Cancer describes a group of diseases characterized by the uncontrolled growth and spread of abnormal cells . It is tERCC2), also called xeroderma pigmentosum group D (XPD), plays important roles in the nucleotide excision repair (NER) pathway  OR “xeroderma pigmentosum group d protein” [All Fields] OR “ercc2” [All Fields]) AND Asp312Asn [All Fields] AND  detailed data available for calculating ORs and the corresponding 95% CIs. Studies were excluded if they: (I) had no control population; (II) were review articles or previous meta-analyses; (III) contained insufficient or duplicate data; or (IV) had no full text available.The following inclusion criteria were set and reviewed by two independent investigators: (I) case-control study; (II) evaluation of the Two authors performed data extraction independently. For all publications, the following data were extracted: first author, year of publication, ethnicity of the population, country, source of cases and controls, cancer site, genotyping method, and number of cases and controls.To evaluate whether our meta-analysis had sufficient sample size to reach firm conclusions about the effect of interventions , TSA wasERCC2 Asp312Asn and cancer risks. In our included studies, no clear models had been chosen; thus, the following genetic models were used: homozygote comparison (Asn/Asn vs. Asp/Asp), heterozygote comparison (Asp/Asn vs. Asp/Asp), recessive model (Asn/Asn vs. Asp/Asp+Asp/Asn), and dominant model (Asn/Asn+Asp/Asn vs. Asp/Asp). The statistical heterogeneity assumption was evaluated using I2 statistics to quantify any inconsistency arising from inter-research variability that was derived from heterogeneity instead of random chance [2 value from 0-25% indicates low heterogeneity, 25-50% moderate heterogeneity and ≥50% high heterogeneity [2< 50%, we used a fixed effect model and when I2 ≥50%, we performed a random effect model [The primary objective of our meta-analysis was to calculate ORs and their 95% CIs to evaluate the association between m chance . An I2 vogeneity . Two modogeneity . When I2ct model , 125. Wect model . Potentict model . Notablyct model . All sta