C20orf54 rs13042395 polymorphism modify the risk of esophageal squamous cell carcinoma (ESCC) and gastric cardia adenocarcinomas (GCA) in common population. We conducted a systematic literature review and evaluated the quality of included studies based on Newcastle-Ottawa Scale (NOS). Pooled odds ratios (ORs) and corresponding 95% confidence intervals (95%CIs) were calculated to estimate the strengths of the associations. 9 articles (10 studies) were identified for synthesis analyses. Overall, the results indicated that the C20orf54 rs13042395 genotype was subtly decrease the risk of ESCC  and the rs13042395 polymorphism was associated with a decreased risk of GCA . The subsets were divided by smoking and drinking status, but none of the genetic comparisons reached statistical significance. Subgroup analysis was also stratified by body mass index (BMI), rs13042395 polymorphism was significantly associated with a subtly decreased cancer risk in under-weight group and normal group, but no association was observed in over-weight group. In conclusion, C20orf54 rs13042395 polymorphism was significantly associated with decreased ESCC and GCA risk especially for the subjects with under-weight or normal.This study aimed to determine whether  Helicobacter pylori (H.pylori) infection is a well-established risk factor for Gastric Cardia Adenocarcinoma (GCA) and has been labeled as a definite human carcinogen by International Agency of Research on Cancer (IARC)6Esophageal cancer and gastric cancer cause more than 400,000 and 700,000 deaths each year, respectively, and represent the sixth and second leading causes of cancer-related death worldwide2In 2010, a large-scale genome-wide association study (GWAS) reported that a new and notable susceptibility locus rs13042395) located in 5’ flanking region of chromosome 20 open reading frame 54 (C20orf54), it encodes riboflavin transporter 2 protein (RFT2) that was newly identified to play an important role in esophageal and carcinogenesis by modulating riboflavin uptake5 located1012C20orf54, it is widely shared that functional genetic variations in the C20orf54 may contribute to the development of ESCC and GCA. The objective of the present study was to quantitatively assess the association between C20orf54 rs13042395 polymorphism and risk of ESCC and/or GCA.For C20orf54 rs13042395 genotype and risk of ESCC and GCA, the results were inconsistent. On the basis of the biological and pathologic significance of The selection process for relevant studies and a flow diagram are shown in 101516171819202122The methodological quality assessment for included studies was summarized in For all of 12 data sets, the frequencies of risk T allele in rs13042395 are presented in C20orf54 rs13042395 polymorphism and the susceptibility to ESCC and GCA is presented in P < 0.01; CT vs. CC: OR = 0.94, 95% CI = 0.88–0.99, P = 0.04; TT vs. CC: OR = 0.91, 95% CI = 0.83–0.99, P = 0.04; CT + TT vs. CC: OR = 0.94, 95% CI = 0.89–0.99, P = 0.01) except recessive model   .C20orf54 rs13042395 polymorphism and ESCC risk was explored in eight studies. The results indicated that the C20orf54 rs13042395 genotype subtly decreased the risk of ESCC, as revealed by the allele genetic model   (The association between  = 0.02) . GCA was < 0.01) .P = 0.73; CT vs. CC: OR = 0.90, 95% CI = 0.66–1.22, P = 0.49; TT vs. CC: OR = 1.00, 95% CI = 0.79–1.27, P = 0.99; CT + TT vs. CC: OR = 0.96, 95% CI = 0.85–1.08, P = 0.48; TT vs. CT + CC: OR = 1.06, 95% CI = 0.84–1.33, P = 0.62) and never smokers   .P = 0.19; CT vs. CC: OR = 0.83, 95% CI = 0.56–1.22, P = 0.34; TT vs. CC: OR = 0.99, 95% CI = 0.75–1.31, P = 0.96; CT + TT vs. CC: OR = 0.95, 95% CI = 0.82–1.10, P = 0.51; TT vs. CT + CC: OR = 1.13, 95% CI = 0.76–1.67, P = 0.54) and never drinkers   .C20orf54 rs13042395 polymorphism was significantly associated with a subtly decreased cancer risk in under-weight group  and normal weight group , but no association was observed in over weight group . One-way sensitivity analyses were performed to assess the influence of the results by the systematic omission of the individual studies from the analyses. The dataset showed that the corresponding pooled ORs were not materially altered, indicating that our results were statistically robust (data not shown).Our data sets indicated that there was no significant heterogeneity between studies among all comparisons in the overall analysis (There was no evidence for publication bias using either Begg’s rank correction. Begg’s funnel plot and Egger’s linear regression test were performed to assess the publication bias of the quantitative synthesis literature. The shape of the funnel plot (Begg’s rank correction) did not reveal any evidence of obvious asymmetry , and no C20orf54 rs13042395 polymorphism and cancer risk (ESCC and/or GCA) were inconclusive. The present study is considered to be the first quantitative meta-analysis concerning the effect of C20orf54 rs13042395 polymorphism on risks of ESCC and GCA and specific stratified analysis . By analyzing the data that extracted from 10 published studies, we revealed that C20orf54 rs13042395 polymorphism might be associated with decreased ESCC and GCA risk especially for under-weight and normal weight groups.Results from previous individual published studies investigating the associations between C20orf54 was significantly associated with ESCC and GCA risk in the GWAS among Chinese population2The genetic basis of ESCC and GCA between a large number of SNPs and disease predisposition has been explored, and the rs13042395 in C20orf54 genotypes modulated the risk of ESCC in smokers, drinkers, or in individuals with a negative family historyC20orf54 rs13042395 polymorphism on the risk of ESCC and/or GCA. However, on this point, our meta-analysis obtained the consistent conclusions came up with Wang et al.ESCC and GCA are complex diseases likely resulting from multiple interacting genetic polymorphisms and gene-environment interactions. Both in the western countries and Asian especially in China, heavy smoking and alcohol consumption were identified as the main environmental risk factors for ESCC and GCA24C20orf54 rs13042395 T allele significantly decreased the risk of ESCC and/or GCA in the subjects with BMI less than 24 especially between 18.5 to 24. Overweight and obesity have been consistently related to gastric and esophageal adenocarcinoma, but not to squamous cell carcinoma262730In the present study, the The following limitations should be acknowledged in our studies. First, the present meta-analysis only included design of case-control studies, some of which were hospital based studies. Thus, the controls may not reflect the representative element of the source population. Second, although all eligible studies were summarized, the relatively small study number may lead to reduced statistical power when stratified according to the cancer type, ethnicity, smoking status, drinking status and BMI. Third, the pooled datasets without excluding the studies with inefficient points based on NOS. In addition, Large-scale studies will be needed for high-risk population screening, individualized prevention, treatment and exposure rating in the future.C20orf54 rs13042395may be associated with a significantly decreased risk of ESCC and GCA, especially for the subjects with BMI less than 24 particularly between 18.5 to 24. Notably, based on the well-designed studies at multicenters with large sample size will be needed for further validate our results.In summary, current data suggest that C20orf54”, “RFT2” and “rs13042395” for both case–control and cohort studies, which evaluated the association between C20orf54 rs13042395 polymorphism and the risk of ESCC and/or GCA . The search was limited to papers published in English or Chinese language. In addition, Reference lists of retrieved articles were examined manually to further identify potentially relevant studies.We comprehensively identified studies through searching PubMed, Embase, Web of Science, Chinese National Knowledge Infrastructure (CNKI) and Wanfang database using terms “C20orf54 rs13042395 and ESCC or GCA; (ii) sufficient allele or genotype data for estimating an odds ratio (OR) with corresponding 95% confidence intervals (95%CIs); (iii) genotype distribution of control groups must be in accordance with the assumptions of Hardy-Weinberg equilibrium (HWE). Case-control studies based on the esophageal adenocarcinomas and/or gastric non-cardia adenocarcinoma were excluded. In case of redundant publications, only the studies with the largest sample size and/or latest published date were included.Studies were included in the analysis if following criteria were met: (i) based on case-control studies (including cohort studies and GWASs) examined the associations between the C20orf54 rs13042395 polymorphism and quality control or not, characteristics of cases and controls. If discrepancies existed, consensus would be finally reached on discussion.Two independent authors (Fujiao Duan and Shuli Cui) extracted the data from the eligible publications. Data for analyses, including first author, publication year, study design, ethnicity, cancer type, source of control, detection methods of We assessed quality of included studies by a modified checklist based on the Newcastle–Ottawa Scale (NOS)We utilized RevMan 5.0  and STATA 12.0  to perform all the statistical analysis.C20orf54 rs13042395 polymorphism and cancer risk by the pooled ORs with corresponding to 95%CIs. The stratified analysis was conducted by ethnicity , smoking status , drinking status  and BMI .RevMan 5.0 was used to estimate the association between χ2) of heterogeneity and the inconsistency index (I2) between each individual study. By heterogeneity test, if P-value for heterogeneity test (Pheterogeneity) < 0.05 or I2 > 50%, the sources of heterogeneity would be used for meta regression in STATA 12.0Pheterogeneity. If Pheterogeneity ≥ 0.05, we used the fixed effect model (the Mantel-Haenszel method)P < 0.05 was considered significant.Heterogeneity was explored by the chi-squared test How to cite this article: Duan, F. et al. Esophageal Squamous Cell Carcinoma and Gastric Cardia Adenocarcinoma Shared Susceptibility Locus in C20orf54: Evidence from Published Studies. Sci. Rep.5, 11961; doi: 10.1038/srep11961 (2015).


Exo1) gene is a key component of mismatch repair (MMR) by resecting the damaged strand, which is the only exonuclease involved in the human MMR system. The gene product is a member of the RAD2 nuclease family and functions in DNA replication, repair and recombination. However, whether Exo1 is required to activate MMR-dependent DNA damage response (DDR) remains unknown, the conclusions of the Exo1 polymorphisms on cancer susceptibility studies were not consistent. We carried out a meta-analysis of 7 case-control studies to clarify the association between the Exo1 K589E polymorphism and cancer risk. Overall,a significant association of the Exo1 K589E polymorphism with cancer risk in all genetic models . In the stratified analysis by ethnicity, significantly increased risk was observed in Asian population . Subgroup analysis based on smoking suggested Exo1 K589E polymorphism conferred significant risk among smokers , but not in non-smokers . In conclusion, Exo1 K589E Lys allele may be used as a novel biomarker for cancer susceptibility, particularly in smokers.The exonuclease1 ( Cancer is the leading cause of death in economically developed countries and the second leading cause of death in developing countries Individual variation in genetic backgrounds can in turn result in different consequences following the environmental exposure and may ultimately determine cancer risk. DNA repair genes form a complex network that protect the genome's integrity from endogenous and exogenous damage Exo1) gene, located at chromosome 1q42–43, contains one untranslated exon followed by 13 coding exons and encodes an 846 amino acid protein Exo1 is a key component of mismatch repair (MMR) by resecting the damaged strand, however, whether Exo1 is required to MMR-dependent DNA damage response (DDR) remains unknown Exo1 polymorphisms on cancer susceptibility studies remain inconsistent, which is partially attributed to the heterogeneity of the cancer subtype, small sample size, and ethnicity of the patients.The exonuclease1 (Exo1 (dbSNP ID: rs 1047840), resulting in the substitution of a glutamic acid  residue (GAG) by lysine  residue (AAG)  in the exonic splicing enhancer (ESE), has been suggested to influence the products of Exo1 mRNA. To further determine whether there is an association of the Exo1 K589E with the risk for developing cancer, a comprehensive review and analysis of published data from different studies is needed.A guanine (G)/adenine (A) common single nucleotide polymorphism (SNP) at first position of codon 589 in exon 13 of Exo1 K589E polymorphisms and cancer susceptibility.In the present study, we have extensively reviewed literature and performed a meta-analysis based on all eligible case-control published data to evaluate the association between Exo1 polymorphism’, and ‘cancer’. The full text of the candidate articles were examined carefully to determine whether they accorded with the inclusion criteria for the meta-analysis. The inclusion criteria were as follows: 1) about the Exo1 K589E polymorphism and cancer risk, 2) from a case-control designed study, 3) sufficient published data for estimating an odds ratio (OR) with 95% confidence interval (CI), and 4) genotype frequencies available.A comprehensive literature search was conducted using the PubMed, Springer, Elsevier, CNKI (Chinese), and Wanfang (Chinese) Digital Dissertations Databases for relevant articles published in English and Chinese up to December 2013 with key words ‘K589E/rs1047840’, ‘The studies, in which the genotype of controls for a certain polymorphism was not consistent with Hardy-Weinberg equilibrium (HWE) were excluded from the analysis of this polymorphism.Data were extracted independently by two investigators. For conflicting evaluations, an agreement was reached following discussion. If they could not reach a consensus, the third investigator was consulted to resolve the dispute, and a final decision was made by vote.The following variables were extracted from each study if available: first author's name, publication year, cancer type, country of origin, ethnicity, study design, genotype distributions, and HWE of controls, respectively. Different ethnicity descents were categorized as Asian or Caucasian. Study design was stratified into hospital-based study and population-based study. If original genotype frequency data were unavailable in relevant articles, a request for additional data was sent to the corresponding author.Exo1 K589E polymorphism were calculated directly from the data given in the eligible studies. OR corresponding to 95%CI was used to assess the strength of association between Exo1 K589E polymorphism and cancer. The pooled ORs were performed for allelic comparison (Lys vs Glu), heterozygote comparison (Glu/Lys vs Glu/Glu) and homozygote comparison (Lys/Lys vs Glu/Glu), dominant model (Lys/Lys+Glu/Lys vs Glu/Glu), recessive model (Glu/Glu vs Glu/Lys + Lys/Lys), respectively. Furthermore, studies were stratified according to ethnicity  and smoking status.The analyses were conducted in Review Manager 5.0. The risks (ORs) of cancer associated with 2χ test with 1 degree of freedom.We assessed the departure from the HWE for the control group in each study using Pearson's goodness-of-fit 2χ based Q-test among the studies P<0.05. In case of no significant heterogeneity, point estimates and 95%CI was estimated using the fixed effect model (Mantel-Haenszel), otherwise, random effects model (DerSimonian Laird) was employed http://www.stata.com).Heterogeneity in meta-analysis refers to the variation in study outcomes between different studies. Between-study heterogeneity was evaluated with a To assess the stability of the results, one-way sensitivity analyses were performed to assess the stability of the results, in which a single study in the meta-analysis was deleted each time to reflect the influence of the individual data set to the pooled OR. The publication bias was diagnosed by using inverted funnel plots, Begg's test and the Egger's test by Stata 12.0.P<0.05.Statistical tests performed in the present analysis were considered significant whenever the corresponding null-hypothesis probability was A total of 8 publications met the inclusion criteria Exo1 K589E polymorphism and cancer susceptibility, stratified analyses by ethnicity and smoking status were performed. When the Q-test of heterogeneity was not significant, we conducted analyses using the fixed effect models. The random effect models were conducted when we detected significant between-study heterogeneity.The main results of this meta-analysis and the heterogeneity test were shown in Exo1 K589E polymorphism and cancer risk in all genetic models .In the overall analysis, we found a significant association between Exo1 K589E polymorphism was significantly linked to cancer risk . In Caucasian population, Exo1 K589E polymorphism was significantly associated with an increased risk in the allelic contrast, homozygote comparison and recessive model .Further stratification analysis by ethnicity, the results showed that cer risk . OverallExo1 K589E polymorphism was significantly associated with an increased cancer risk in smokers , but no association was observed in non-smokers .Subgroup analysis was also stratified by smoking status. Begg's funnel plot and Egger's test were performed to assess the publication bias of the currently available literature. The shape of the funnel plots did not reveal any evidence of obvious asymmetry in all comparison models . Then, tA single study involved in the meta-analysis was deleted each time to reflect the influence of the individual data-set to the pooled ORs, and the corresponding pooled ORs were not materially altered, indicating that our results were statistically robust (data not shown).Exo1 is a member of the RAD2 family of nucleases and possesses 5′ to 3′ double-stranded DNA (dsDNA) exonuclease and 5′–flap endonuclease activities and functions in a number of important cellular pathways including DNA repair, replication, recombination, and telomere integrity Exo1 is the only exonuclease involved in the human MMR system, one of the major roles is the MMR system which is responsible for correcting mismatches between bases and small insertion or deletion loops NQO1, CYP1A1, ERCC4, EXO1, MSH2, XRCC1 and hOGG1 have been identified, only some of them have been extensively investigated in epidemiological studies Exo1 gene.Exo1 K589E of cancer from 7 studies. The pooled results revealed that Exo1 K589E Lys allele was associated with an increased risk for developing cancer. Among Asian population, Exo1 K589E polymorphism was significantly associated with an increased cancer risk in all genetic models but not in the Caucasian population, this suggested that a possible ethnic difference in the genetic background. Subgroup analysis was stratified by smoking status, Exo1 K589E polymorphism was significantly associated with an increased cancer risk in smokers, but no significant association was observed in non-smokers. The reasonable explanation is cigarette smoking, a well-known origin of DNA damage, releases many DNA damage inducers to respiratory system and causes DNA damages to the cells. Therefore, people who have high-risk genetic variant, such as the Lys allele of K589E, and also smoking habits, the combined effect of genetic and environmental factors would synergistically increase their cancer susceptibilities.In the present study, we were first analyzed the association of Although meta-analysis is robust, our study still has some limitations. Firstly, lacking sufficient eligible studies limited our further stratified analysis on types of cancer. Secondly, for each selected case-control study, our results were based on unadjusted estimates, whereas a more precise analysis could be performed if individual data were available. Thirdly, lack of the original data of the reviewed studies limited our further evaluation of potential interactions, because the interactions between gene-to-gene and gene-to-environment may modulate cancer risk. Fourthly, although all eligible studies were summarized, the relatively small sample size of studies may lead to reduced statistical power when stratified according to the ethnicity and smoking status.Exo1 K589E polymorphism was significantly associated with increased risk of cancer, especially in smokers. However, further well-designed studies in large cohort of different ethnic origins and cancer types are needed before the application of Exo1 K589E polymorphism as cancer biomarker in clinical settings and early cancer detection.In summary, this meta-analysis suggested that the Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.


GST genes may exemplify molecular connectors between ATDILI and SCZ. However, association studies of GSTM1/T1 polymorphisms with these two diseases have yielded conflicting results. After searching case-control association studies in PubMed, ISI Web of Science, EMBASE, Chinese National Knowledge Infrastructure (CNKI), and Chinese BioMedical Literature Database, we performed meta-analyses across a total of 20 published association studies on 3146 subjects for the association of GSTM1 and ATDILI, 2587 for the GSTT1-ATDILI association, 2283 for GSTM1-SCZ and 1116 for GSTT1-SCZ to test the associations of GSTM1/T1 polymorphisms with ATDILI and SCZ. The GSTM1 present genotype was significantly associated with decreased risks of ATDILI (risk ratio(RR): 0.81, 95% confidence interval (CI): 0.75–0.88, P < 0.0001) and SCZ  according to the fixed-effect model, while the GSTT1 present genotype was significantly associated only with a high risk of SCZ  according to both the random- and fixed-effect models, but not with ATDILI (P = 0.82) according to the fixed-effect model. Moreover, these significant results were supported with moderate evidence according to the Venice criteria. These results indicate that GSTM1 represents a genetic connection between ATDILI and SCZ, and suggest that ATDILI and SCZ may be co-occurring for the subjects with GSTM1 null genotype.Anti-tuberculosis drugs have some adverse effects such as anti-tuberculosis drug-induced liver injury (ATDILI) and mental disorders. The involvement of glutathione S-transferase (GST) genes in pathogenesis of ATDILI or schizophrenia (SCZ) has been reported. Therefore,  GSTs) related disease [Tuberculosis (TB) remains a devastating disease and the major leading cause of death worldwide. In 2011, an estimated 8.7 million new TB cases were reported, and 1.4 million people died from TB –5. ATDIL disease –8.GST genes are hypothesized to play an important role in both ATDILI and SCZ, acting as ‘molecular bridges’ between these two diseases.Schizophrenia (SCZ) is a severe, disabling and chronic mental disorder affecting approximately 1% of the general population ,10. OxidGSTM1) gene, located on chromosome 1p13.3,which codes for cytosolic GST class Mu 1 enzyme, and the glutathione S-transferase theta-1 (GSTT1) gene, located on chromosome 22q11.2, which codes for cytosolic GST class theta 1 enzyme [GSTM1 and GSTT1 have a null mutation consisting of the complete deletion of the respective gene through homologous unequal crossing over [GSTM1 and GSTT1 gene can cause the absence of GST activity, and thus may induce different diseases, including ATDILI and SCZ. However, so far the studies testing the association between GSTM1/T1 polymorphisms and SCZ or ATDILI have produced conflicting results [As phase II detoxification enzymes, GSTs have important protective effects of detoxification of anti-TB drugs’ reactive metabolites and products of oxidative stress through conjugating glutathione with target toxic substances and facilitating their elimination from the body ,15. GSTs1 enzyme ,17. Bothing over –20. Homo results –23.GST genes represent ‘molecular connectors’ between ATDILI and SCZ, we have performed meta-analyses of published studies that test the association between GSTM1/T1 polymorphisms and ATDILI or SCZ.To assess whether the GST’, ‘GSTM’, ‘GSTM1’, ‘GSTT’, or ‘GSTT1’). Furthermore, references of retrieved articles were also reviewed for the literature that they cite.The digital medical databases PubMed, ISI Web of Science, EMBASE, Chinese National Knowledge Infrastructure (CNKI) and Chinese BioMedical Literature Database were searched for studies with publication date up to Nov. 30 2013 using the following keywords:  and (‘Schizophrenia’), respectively, combined with  original case-control association studies with complete data were based on unrelated, randomly selected individuals; 2) cases were TB patients with ATDILI and controls were TB patients without ATDILI, or cases were Schizophrenic patients and controls were healthy subjects; 3) both cases and controls were matched for sex and age; 4) association studies of GSTM1/T1 polymorphisms with SCZ or ATDILI. Other studies, such as: case-only studies, duplications, animal studies, comparisons of laboratory methods, editorials, and review articles were excluded.Data extraction was performed independently by two reviewers using a standardized protocol and reporting form. The discrepancy between the two reviewers was resolved by further discussion with a third party. For overlapping studies, the study with the larger sample size was retained for the meta-analysis. The recorded study characteristics included: 1) first author’s name; 2) publication year; 3) sample ethnicity; 4) number of cases and controls for GSTM1/T1 null and present genotypes; 5) control and case characteristics; 6) sex proportion; 7) mean age of cases and controls; 8) methods used for genotyping; and 9) PCR primers and amplification regions.GSTM1 and GSTT1 polymorphisms and the risks of ATDILI or SCZ were measured using risk ratios (RRs) with corresponding 95% confidence intervals (CIs)[I² statistic was also calculated to quantify the proportion of the total variation due to heterogeneity, and I² >50% was considered to be statistically significant [The strengths of the associations between als (CIs). Pooled als (CIs),26. The als (CIs). Betweenals (CIs). The I² nificant . If the nificant . A sensinificant . All staminor >1,000, grade B for 100≤nminor≤1,000 and grade C for nminor <100. Here, the nminor referred to the total number of cases and controls with the least frequent genotype. To assessment of replication, grade A was given for I²<25%, grade B for 25%≤I²≤50% and grade C for I²>50%. To assess bias protection, any of following criteria were required: 1) clear phenotype definition; 2) high genotyping quality with a low genotyping error rate; 3) no loss of significance when the first published study was excluded; and 4) no evidence of small-study effects based on a Harbord regression test  [The cumulative evidence for the genetic association of GSTM1 and GSTT1 present genotypes with ATDILI/SCZ, respectively, was assessed according to the Venice interim criteria including amount of evidence, replication of results and protection from bias . With re P<0.05) .GSTM1/T1 polymorphisms and respective risk of ATDILI and SCZ, respectively, were identified after firstly screening based on the titles and abstracts of the candidate articles. After the second screening, totally, there were 965 cases and 1881 controls in 13 studies of GSTM1 and ATDILI, 811 cases and 1476 controls in 11 studies of GSTT1 and ATDILI, 1101 cases and 1182 controls in 5 studies of GSTM1 and SCZ, and 568 cases and 548 controls in 3 studies of GSTT1 and SCZ. The detailed characteristics of each study are listed in GSTM1 or GSTT1. The genotype distributions across all studies for cases and controls of ATDILI and SCZ are shown in A flow diagram summarizing the study selection process was shown in GSTM1/T1 polymorphisms and risk of ATDILI is summarized in I2 = 22% for GSTM1, P = 0.37 and I2 = 8% for GSTT1), the fixed-effect model was used to analyze both GSTM1 and GSTT1. The frequencies of both GSTM1 and GSTT1 null genotype were higher among cases than among controls (54.51% vs. 42.42% for GSTM1 and 38.84% vs. 36.92% for GSTT1). This finding indicated that the GSTM1 present genotype was significantly associated with a decreased risk of ATDILI    , whereas = 0.82) .GSTM1 and GSTT1 genotypes and ATDILI . The regression model indicated strong evidence for an association between the race and the effects of the GSTM1 present genotype (P = 0.033). However, although the total sample size power for detecting a significant effect between the GSTM1 present genotype and ATDILI exceeded 99%, the sample size power in the East Asian group exceeded 99%, whereas that in the Caucasian group was only 14.8%. Given the low power value in the Caucasian group, we are unable to definitively conclude on the association between GST genes and ATDILI in the Caucasian populations.In subgroup analyses of the association between d ATDILI , a signiGSTM1/T1 polymorphisms and risk of SCZ are shown in GSTM1 (P = 0.22 and I2 = 41%), the fixed-effect model was used; because mildly significant heterogeneity was observed in analysis of GSTT1 , the random-effect model was used. The GSTM1 null genotype and GSTT1 present genotype frequencies were higher among cases than among controls (56.95% vs. 51.02% for GSTM1 and 23.94% vs. 35.77% for GSTT1). This finding indicated that statistically significant associations between the GSTM1 present genotype and a decreased risk of SCZ  in the fixed-effect model  in the random-effect model . The total sample size power for detecting a significant association between GSTM1 present genotype and SCZ was 83.5%; in contrast, sample size powers were 68.2% for the East Asian group and 32.1% for the Caucasian group.Because only three papers concerning the with SCZ . HoweverP values for testing the overall effects after excluding one study ranged from <0.00001 to <0.0001 and from 0.37 to 0.90 in the GSTM1 and GSTT1 and ATDILI analyses, respectively; the P values for testing overall effects ranged from 0.0006 to 0.03 in the GSTM1 and SCZ analyses. These results indicated that the analyses were stable. However, in GSTT1 and SCZ analysis, the summarized RRs ranged from 1.17 (95%CI: 1.04–1.32) to 1.12 (95%CI: 0.94–1.34) and the P values from 0.01 to 0.20 after excluding a study by Saadat (13); this indicated that this study was the main source of the observed mild heterogeneity. Furthermore, there was no indication of significant publication bias in the overall meta-analysis according to both Egger’s and Begg’s tests .A sensitivity analysis via re-analysis after leaving one study out was performed to assess the effect of individual studies on the overall meta-analysis estimate. The minor for GSTM1 null genotype was 1321, indicating a grade of A. The I2 for the GSTM1 null genotype was 22%, and a grade of A was given. After excluding study by Roy conducted in 2001 [GSTM1 present genotype and ATDILI (P<0.0001), and the Harbord test P value was 0.679. For the meta-analysis of SCZ studies, nminor for the GSTM1 present genotype was 1053, and a grade of A was given. The I2 for the GSTM1 null genotype was 41%, resulting in a grade of B. After excluding the study by Harada conducted in 2001[GSTM1 present genotype and SCZ (P = 0.03), and the Harbord test P value was 0.0716. Therefore, ‘moderate’ cumulative evidence supported significant associations of the GSTM1 present genotype with both ATDILI and SCZ.In this meta-analysis, strict inclusion criteria were used to address phenotype definitions and genotyping quality. For the meta-analysis of ATDILI studies, the nGSTM1 present genotype was significantly associated with decreased risks of ATDILI (P < 0.0001) and SCZ (P = 0.004), whereas the GSTT1 present genotype was only significantly associated with a high risk of SCZ (P = 0.01), but not ATDILI (P = 0.82); these significant results were supported by ‘moderate’ evidence according to the Venice criteria. Because a dependent conclusion may have resulted from the evidence that a number of studies simultaneously tested the associations of both GSTT1 and GSTM1 with single disorder, a Bonferroni correction method was used to correct the P value. After correction, the GSTM1 present genotype remained significantly associated with decreased risks of both ATDILI (P < 0.0001) and SCZ (P = 0.008), and the GSTT1 present genotype remained significantly associated only with a high risk of SCZ (P = 0.02), but not ATDILI (P = 1). These results indicate that GSTM1, rather than GSTT1, may be a ‘molecular connector’ between ATDILI and SCZ.Previously, ATDILI was recognized as the primary adverse effect of anti-TB drugs. Although mental disorders, of which SCZ is an example, have been considered an additional adverse effect of anti-TB drugs , a clearGSTM1 is mainly expressed in the liver and brain [GSTM1 is found to not only detoxify anti-TB drugs’ toxic metabolites generated by CYP2E1 in the liver, but also catalyze the conjugation of glutathione with the aminochrome and the dopa-o-quinone metabolite of oxidized dopamine in the brain [GSTM1 caused by the GSTM1 null genotype can cause not only liver injury but also accumulation of neuro destructive oxidants leading to Schizophrenia [GSTT1 has considerable presence in the brain, it has only trace expression in the liver [GSTT1 can act as a deleterious metabolic activator for halogenated compounds to produce a variety of intermediate materials potentially dangerous for DNA and cells [GSTT1 gene and ATDILI, but indicated that the GSTT1 present genotype increases the risk of SCZ.GSTs, acting as free radical scavengers through glutathione conjugation to reduce target substances’ potential toxicity, have varied tissue-specific expression patterns . GSTM1 ind brain , which ohe brain . In the he brain . Thus, tophrenia . In conthe liver , and it he liver . Furthernd cells . In the I2 were estimated for the heterogeneity test. No significant heterogeneity was observed in the meta-analyses, except for mildly significant heterogeneity in the analysis of GSTT1 with SCZ. Interestingly, in the subgroup analyses of studies concerning GSTM1 and ATDILI, a significant association was found in the East Asian group, but not in the Caucasian group. This may be because of the low power value in the latter group (14.8%), which reduces the ability to detect significant effects. Although race was significant associated with the effects GSTM1 on ATDILI, the biological impact of GSTM1 gene on the risk of ATDILI may be consistent across racial groups, and this significant result may instead be because of the effects of different cultural and living habits among different races, such as: diet [GSTT1 with SCZ and indicated that the study by Saadat greatly affected this analysis; accordingly, these results should be treated with caution. Two meta-analyses have been previously published on GSTM1/T1 and ATDILI: Cai in 2012 and Tang in 2013 [GSTM1 null genotype was significantly associated with an increased risk of ATDILI, a study by Kim was included in those previous meta-analyses, but was excluded from the current analysis[A meta-analysis with a larger sample size is widely considered to provide greater statistical power to identify the effect of a genetic polymorphism on a disease than a single association study . An asseas: diet . A sensi in 2013 ,43. Alth analysis. In Kim’Some potential limitations of our meta-analysis deserve consideration. First, diseases are caused by both genetic and environmental factors; however, we did not include an environmental factor analysis. We attempted to obtain complete information regarding dietary habits and substance abuse, among other factors. However, the collected literature seldom provided data in addition to genotypic data. As a result, the current meta-analysis solely focused on genetic factors. Second, the sample size in the current analysis was limited. Additional association studies focused on the co-morbidities of ATDILI and SCZ, and involving large-scale samples will be required to confirm the current results.GSTM1, rather than GSTT1, may act as a molecular link between ATDILI and SCZ and suggest that ATDILI and SCZ may be concurrent adverse effects of anti-TB drugs in subjects harboring the GSTM1 null genotype.In summary, the current meta-analysis results demonstrate that S1 FigA. the summary of RRs with 95% CIs for GSTM1 present genotype and ATDILI; B. the summary of RRs with 95% CIs for GSTT1 present genotype and ATDILI; C. the summary of RRs with 95% CIs for GSTM1 present genotype and Schizophrenia.(DOCX)Click here for additional data file.S1 Table(XLS)Click here for additional data file.


Consumption of dietary fat has been reported to be associated with gastric cancer risk, but the results of epidemiologic studies remain inconsistent. We conducted a meta-analysis to summarize the evidence regarding the association between dietary fat intake and gastric cancer risk.2 and tau2(between study variance)statistics. Subgroup analysis and publication bias analysis were also performed.A comprehensive search of PubMed and EMBASE was performed to identify observational studies providing quantitative estimates between dietary fat and gastric cancer risk. Random effects model was used to calculate the summary relative risk(SRR) in the highest versus lowest analysis. Categorical dose-response analysis was conducted to quantify the association between dietary fat intake and gastric cancer risk. Heterogeneity among studies was evaluated using IP< 0.001; tau2 = 0.12; I2 = 69.5%, 95% CI: 55%-79%) and 1.08 with a daily increase in total fat intake (20 g/d) . Positive association between saturated fat intake , inverse association between polyunsaturated fat intake  and vegetable fat intake , and no association between monounsaturated fat intake  and animal fat intake  and gastric cancer risk were observed.Twenty-two articles were included in the meta-analysis. The SRR for gastric cancer was 1.18 for individuals with highest intake versus lowest intake of total fat (95% confidence interval [CI]: 0.999–1.39; n = 28; Our results suggest that intake of total fat is potentially positively associated with gastric cancer risk, and specific subtypes of fats account for different effects. However, these findings should be confirmed by further well-designed cohort studieswith detailed dietary assessments and strict control of confounders. Helicobacter pylori infection and dietary factors play important roles in the etiology of gastric cancer[Helicobacter pylori infection was reported to be positively associated with the riskof GNCAbut not ofGCA[Gastric cancer, which account for about 10% of annual cancer-related deaths, remains one of the most common malignant tumors worldwide, especially in developing countries, such as EasternAsia, Eastern Europe, and South America[ic cancer.A distinic cancer.For examnot ofGCA, while onot ofGCA.However,not ofGCA. In spitnot ofGCA–9, whilenot ofGCA. OverallDietary fat has been reported to be associated with various malignant tumors, such as breast cancer, colorectal cancer, pancreatic cancer, and prostate cancer –17.HowevGiven the inconsistent results of previous observational studies, we performed this meta-analysis to summarize the evidence regarding the association between dietary fat intake and gastric cancer risk.A computerized literature search of PubMed and EMBASE databases was conducted to indentify relevant literatures published up to March 2015 by two independent investigators (Han and Jiang). We used mesh words and text words (“gastric neoplasm (cancer)” OR “stomach neoplasm (cancer)”) AND  to identify relevant studies. Reference lists of review articles and retrieved articles were also reviewed for additional relevant publications.1) study designed as case-control or cohortstudy,2) study evaluated the association between intake of dietary fat and gastric cancer risk,3) study reported odds ratio (OR)or relative risk (RR) and 95% confidence intervals(CI)according to the highest versus the lowest intake of dietary fat or data sufficient to calculate them, and4) study was published as a full paper in English. When the studies used the same data series, only the study with most complete results or with largest population was included.The eligibility criteria of the studies were as follows:Two authors (Han and Jiang)independently extracted data from eligible studies using standardized forms . DisagreA 9-star system on the basis of the Newcastle-Ottawa Scale (NOS) was used to assess the study quality.A total All statistical analyses were performed with Stata version 11.0 . In this analysis, we assumed OR to be comparable with the RR. For studies that only reported results for GCA and GNCA, men and women, or intestinal type and diffuse type separately, we considered them as independent data obtained from different studies.If no more than 3 studies reportedthe association between a specific subtype of fat and gastric cancer risk,we did not summary the overall effect of it.2 andtau2(between study variance)statistics[P value <0.05 was considered to be of significant publication bias.Firstly, we conducted a meta-analysis to compare the highest category with the lowest category intake of dietary fat.DerSimonian-Laird random effects model, accounttatistics. Sourcestatistics. Subgroutatistics, 30. AndNext, to assess the dose-response relationship between intake of dietary fat and gastric cancer risk, we conducted a meta-analysis of dose-response categorical data using the method proposed by Greenland and Orsini et al (2-stage GLST in Stata), 32. StuWe identified 8087 articles after searching PubMed and EMBASE database. Titles and abstracts of 614 articles were reviewed after removing duplicate articles, non-original articles, non-human articles, and non-English articles. Full texts of 33 articles were reviewed after excluding 581 articles with no reports of the association between dietary fat intake and gastric cancer risk. Out of these33 articles, 6 articles were excluded because they shared the same population with another 2 articles, 5 articCharacteristics of the included studies are shown in The quality scores of included studies ranged from 4 to 9 based on the NOSscore system. The one cohort study and 12 case-control studies were considered to be high quality studies(NOSscore≥7), whereas 9 other case-control studies were considered to be low quality studies (NOSscore <7).P< 0.001; tau2 = 0.12; I2 = 69.5%, 95% CI: 55%-79%).Twenty-one articles with 28 studies have investigated the association between total fat intake and gastric cancer risk. The SRR for the highest compared with lowest analysis was 1.18 , studies located in North America, population-based case-control studies , studies using HHHQ , and studies with a small sample size or low quality. As shown in Subgroup analyses were conducted to examine thestability of the pooled RRs. The results indicated that a positive association between the total fat intake and gastric cancer risk was observed in studies of GNCA patients . No evidence of publication bias was observed in the subgroup published after 2000, as tested using Egger’s (P = 0.67) and Begg’s (P = 1.00) tests. But obvious publication bias was observed in the subgroup published before 2000, as tested using Egger’s (P = 0.02) and Begg’s (P = 0.27) tests. The results showed that publication bias was more likely in older studies, which was in agreement with previous study [We also performed subgroup analyses according to different publications years (before 2000 and after 2000) to investigate the association between total fat intake and gastric cancer risk. The results showed no significant difference between two groups  to 1.22 . Therefore,the non-significant association between total fat intake and gastric cancer risk should be interpreted with caution. No evidence of publication bias was observed in the funnel plot, as tested using Egger’s (P = 0.43) and Begg’s (P = 0.29) tests tests.P = 0.09; tau2 = 0.002;I2 = 46.8%, 95% CI: 0%-79%).However, the increased risk of total fat increase intake per 20 g/d was only found in 4 case-control studiesbut not in 2 cohort studies .There was no evident non-linear association between total fat intake and gastric cancer risk and 3case-control articles (4studies) providing the distribution of cases and controls (person-year)were eligible for the dose-response analysis. The results showed that the SRR per 20 g/d total fat increase intake was 1.08 (95%CI: 1.02–1.14), with moderate heterogeneity .Fourteen articles with 18 studies have investigated the association between saturated fat intake and gastric cancer risk. The SRR for the highest compared with lowest analysis was1.31 (95% CI: 1.09–1.58), indicating a significant positive association. HoweverP = 0.44; I2 = 0.0%), studies located in North America , population-based case-control studies , studies with a small sample size , and high quality studies , but not in other subgroups  and Begg’s (P = 0.34) tests. Unfortunately, only 3studies (from 2 articles) were eligible for the dose-response analysis. Thus, we did not perform dose-response analysis in this group.Sensitivity analysis showed that no individual study could change the summary positive association between saturated fat intake and gastric cancer risk. There was no evidence of publication bias, as tested using Egger’s, indicating a significant inverse association. However, significant heterogeneity was detected  or GNCA patients only , studies located in North America , cohort studies , population-based case-control studies , and studies with a large sample size . In other subgroups, the summary results were consistent with the overall results of all studies  and Begg’s (P = 0.06) tests. We did not perform dose-response analysis in this group due to very few eligible studies.No individual study could change the summary results in the sensitivity analyses. No significant evidence of publication bias was detected, as tested using Egger’s.Elevenarticles with 14 studies have investigated the association between monounsaturated fat intake and gastric cancer risk. The SRR for the highest compared with lowest analysis was1.00 (95% CI: 0.79–1.25), with significant heterogeneity among studies (P = 0.67) and Begg’s (P = 0.74) tests. Dose-response analysis was not performed due to too few eligible studies.In subgroup analyses, the results were fairly consistent with the overall summary estimates when the analyses were restricted to any subgroup, indicatP = 0.13;tau2 = 0.02; I2 = 42.0%, 95% CI: 0%-70%) and 0.55 , respectively. The results indicated that consumption of vegetable fat but not animal fat could reduce the risk of gastric cancer. Sensitivity analyses showed SRRs were not markedly modified by any individual study in two groups.Out of the included studies, 6studies (from 5articles)and 4studies have investigated the associations between intake of animal fat and vegetable fat and gastric cancer risk, respectively. The SRRs for the highest compared with lowest analysis were1.10 , 67–69. The major monounsaturated fatty acid found in human diet isoleic acid. Mediterranean diet, characterized by the high consumption of olive oil, rich in oleic acid, has been traditionally linked to a protective effect on cancers . In a coOpposite to our findings of protective effect of polyunsaturated fat on gastric cancer, polyunsaturated fat was considered to increase the risk of breast cancer and had no effect on colorectal cancer, 68. It In this meta-analysis, we observed an inverse association between vegetable fat intake and gastric cancer risk and no association for animal fat.However, it has been shown that for small meta-analyses homogeneity is often assumed when this is not the case and existing heterogeneity is not detected . In addiOur meta-analysis had several strengths. As far as we known, it is the first meta-analysis focused on the association between dietary fat intake and gastric cancer risk. Our meta-analysis included a large number of studies and more than 9,000 cases, and over 500,000 non-cases in this analysis. Thus we had adequate statistical power to clarify the association between dietary fat intake and gastric cancer risk. We have also quantified the association between intake of dietary fat and gastric cancer risk by carrying out dose-response analyses and the results were generally consistent with the highest versus lowest analysis.Helicobacter pylori infection were not adjusted in most studies. Thus, the possibility of residual confounders as potential explanation for the observed association between dietary fat intake and gastric cancer risk could not be excluded. Moreover, it could not be excluded that other nutrients or dietary components that were correlated with dietary fat might be responsible for the observed associations. Fourth, most studies used food frequency questionnaires to assess dietary fat intake. However, as reported before, measurement errors associated with the use of food frequency questionnaires may obscure associations between dietary fat intake and gastric cancer risk[This meta-analysis has several limitations. First, only one cohort study was included and the rest 21 studies were case-control studies, which may introduce recall bias and selection bias, especially the hospital-based case-control studies.The associations between intake of polyunsaturated fat and saturated fat and gastric cancer risk were not statistically significant in the cohort study but are statistically significant in the case-control studies. More cohort studies are needed to confirm the association in this cohort study. Second, most of included case-control studies did not report the doses or number of cases and controls for each category. Therefore, the results of dose-response analysis were not credible because too few eligible studies were available for the dose-response meta-analysis. Future studies should report the complete results with standardized dietary fat categories and number of cases and control in order to assess the dose-response relationship. Third, a majority of studies only adjusted forage, sex, and total energy intake. Some other known risk factors for gastric cancer such as family history, BMI, and ncer risk. Fifth, ncer risk.Finally,In conclusion, results from this meta-analysis indicate a possible significant positive association between total fat intake and gastric cancer risk. Consumption of saturated fat is positively associated with gastric cancer risk. Consumption of polyunsaturated fat and vegetable fat is inversely associated with gastric cancer risk. Consumption of monounsaturated fat and animal fat is not associated with gastric cancer risk. Caution is needed in interpreting these associations because the results of subgroups analyses were not always consistent. Additional well-designed cohort studies with detailed dietary assessment sand strict control of confounders are needed to confirm the associations between dietary fat intake and gastric cancer risk.S1 Checklist(DOC)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.


We identified 21 studies examining 19 variants in 11 genes. The most commonly studied variant was a single-nucleotide polymorphism in ANKK1 . Fixed- and random-effects meta-analyses of this variant  revealed that striatal BP was significantly and robustly lower among carriers of the minor allele (Lys713) relative to major allele homozygotes. The weighted standardized mean difference was −0.57 under the fixed-effect model , P=0.0002). The normal relationship between rs1800497 and BP was not apparent among subjects with neuropsychiatric diseases. Significant associations with baseline striatal D2 receptor BP have been reported for four DRD2 variants  and a PER2 repeat polymorphism, but none have yet been tested in more than two independent samples. Our findings resolve apparent discrepancies in the literature and establish that rs1800497 robustly influences striatal D2 receptor availability. This genetic variant is likely to contribute to important individual differences in human striatal function, neuropsychiatric disease risk and pharmacological response.The D2 dopamine receptor mediates neuropsychiatric symptoms and is a target of pharmacotherapy. Inter-individual variation of D2 receptor density is thought to influence disease risk and pharmacological response. Numerous molecular imaging studies have tested whether common genetic variants influence D2 receptor binding potential (BP) in humans, but demonstration of robust effects has been limited by small sample sizes. We performed a systematic search of published human  D2 receptors mediate an array of fundamental brain functions, including reward behavior, regulation of movement, learning and memory, and attention. The D2 receptor is also important clinically as a target of pharmacotherapy for psychosis, parkinsonism, brain injuries and restless legs syndrome.in vivo imaging studies, with typical coefficients of variation in the range of 20–60%.7 Individual differences in D2 receptor expression are hypothesized to contribute to differences in motivated behaviors and risk of related neuropsychiatric disorders. For example, striatal D2 receptor manipulation in rodents alters motivated behaviors,8 and human studies have consistently demonstrated abnormally low striatal D2 receptor levels among individuals with addictions.9 The factors that underlie individual differences in D2 receptor expression in humans are not yet defined, but presumably both environmental and genetic factors are at work.Substantial inter-individual variation in D2 receptor expression has been observed in human postmortem and in vivo molecular imaging (positron emission tomography (PET) or single-photon emission computed tomography (SPECT)) to test whether common genetic variants influence D2 receptor binding potential (BP) in humans. However, consistent and robust genetic effects have not yet been demonstrated, and the magnitude of genetic effects is unknown. For example, among seven published studies of the single-nucleotide polymorphism (SNP) rs1800497, four11 reported a significant association and three13 found no significant effect on striatal BP. There are several potential causes of such discrepancies. A genetic variant may have no true underlying effect (type I error), the true effects may be different in different studies (heterogeneity), or a true effect may be present but not consistently detected due to the low power of individual studies (type II error). This latter possibility may be particularly relevant to molecular imaging studies as their cost and complexity often necessitate small sample sizes.To address the role of genetic factors, numerous studies have used 14 Similarly, meta-analyses of the variable repeat polymorphism in the promoter of the serotonin transporter and amygdala activation have indicated a much smaller effect size than initially reported.16 Our objective here was to perform a review and quantitative analysis of all published studies that tested genetic polymorphisms and measured human in vivo D2 receptor BP, thereby establishing associations that are known with confidence.Recent work has highlighted the importance of resolving such apparent discrepancies in the genetics-imaging literature. For example, despite compelling preclinical evidence and numerous human imaging studies, a recent meta-analysis of the brain-derived neurotrophic factor Val66Met polymorphism found no significant effect on hippocampal volume.We performed a systematic search of PubMed in May 2015 using the following search string: ‘(variant* OR polymorphism* OR genetic OR gene) AND (pet OR positron OR emission OR spect OR photon OR tomography) AND (dopamine OR d2 OR d3) AND receptor.' The titles and abstracts of all search results were reviewed to identify studies that examined genetic polymorphisms and used human D2 molecular imaging (PET or SPECT). Full text was retrieved for all potentially relevant articles. The references of full text articles were scanned to identify additional sources missed by the original search.http://cran.r-project.org/web/packages/metafor) and ‘meta' package  with RStudio (ver. 0.97.551) within the R statistical computing environment . The effect size was represented by the standardized mean difference (SMD), that is, the mean BP for one genotype group minus mean BP for the second genotype group, divided by Hedges' pooled s.d. Inverse-variance weighting was used to compute the pooled SMD. Fixed-effect and random-effects models were estimated. Forest plots and funnel plots were inspected for outliers and bias. Sensitivity analyses used the leave-one-out method to test for undue influence of single studies. When Cochran's Q-test for heterogeneity indicated nonsignificant heterogeneity across studies (I2<0.5 and P>0.10 as suggested17), we adopted the effect size from the fixed-effect model as our final estimate. Only BP values from anatomically defined regions of interest were used in meta-analyses. Because BP estimates from different brain regions within the same sample are highly correlated, and valid meta-analysis requires independent samples, a single estimate from each genotype group was derived (pooled mean and s.d. with inverse-variance weighting) in cases where BP estimates from multiple anatomical regions were reported. Bilateral striatum was preferred and was used for all studies except Savitz et al.11 who reported a middle caudate region of interest. To avoid the confounding effects of disease and treatment, medication-free healthy control subjects were analyzed separately from subjects with neuropsychiatric disorders. Studies with different demographic characteristics (sex and age distribution) and imaging methods  were first analyzed together based on the assumption that these study-level variables have effects on binding measures that are similar in the two genotype groups. Although the small number of studies severely limited our power to detect statistical effects of these study-level variables on the estimated genetic effect, we did explore these potential moderators with fixed-effect meta-regression and leave-one-out sensitivity analysis.Meta-analysis was performed when possible for genetic variants tested in multiple independent samples. We used the ‘metafor' package  except rs1079597 and rs1076560 (r2=0.87).We identified 21 studies that examined genetic polymorphisms and used human D2 molecular imaging . The 21 ANKK1) gene and 10 kb downstream from the dopamine D2 receptor (DRD2) gene. This missense mutation (C>T) is predicted to change a glutamate residue to a lysine residue at position 713 (Glu713Lys) within the 11th ankyrin repeat of the ANKK1 protein.33 This variant is often referred to as the ‘Taq1A' polymorphism  with the minor Lys713 allele denoted ‘A1' and the major Glu713 allele denoted ‘A2'.34The most commonly studied variant was rs1800497, a SNP within the ankyrin repeat and kinase domain containing 1  under the fixed-effect model and −0.56 , P=0.003) under the random-effects model. Sensitivity analysis showed that the results were robust when individual studies were omitted  indicating that no single study unduly influenced the results. We found no evidence of heterogeneity across studies =4, I2=0.29, P=0.22). Likewise, a funnel plot did not suggest heterogeneity or bias (data not shown). Thus, our results are consistent with a single underlying effect of the Taq1A polymorphism across studies.Five of the eight Taq1A studies were compatible with meta-analysis. One article was not compatible because baseline BP values were not reported.ozygotes . The weiβ=0.024, s.e.m.=0.011, Q=4.4, df=1, P=0.035) with stronger genetic effects evident in studies with a greater proportion of females. Meta-regression revealed no significant effects of imaging modality (P=0.48), brain region (P=0.27) or mean subject age (P=0.38). Finally, fixed-effect analyses that excluded the SPECT study5 or the study that quantified a middle caudate region11 produced similar estimates of effect size .Meta-regression and sensitivity analyses were used to explore the potential influence of study imaging modality (PET versus SPECT), brain region of interest, sex distribution and mean age on the Taq1A findings. We considered these exploratory analyses, as our power to detect effects of study-level variables across five studies was very limited. Meta-regression suggested that sex distribution may be a moderator of the Taq1A effect  among A1-allele carriers relative to A2 homozygotes .Three of the Taq1A ozygotes . The effin vivo imaging studies, two studies of postmortem brain have reported binding density (Bmax) versus Taq1A status,3 so we extracted effect sizes from these studies as well. Striatal binding was generally lower among healthy A1-allele carriers relative to A2 homozygotes , is sometimes called the ‘Taq1B' polymorphism, with the minor allele denoted ‘B1' and the major allele denoted ‘B2'.36 The first study5 reported no significant effect of this polymorphism on D2 binding but the second study4 reported significantly lower striatal BP among healthy B1-allele carriers relative to healthy B2 homozygotes. Although this apparent discrepancy has been interpreted as a replication failure,4 the effect sizes are actually consistent with each other ; and mean=−0.72, 95% CI=, respectively) and the weighted mean effect size from the two samples is −0.58 , P=0.008). Thus, the Taq1A and Taq1B polymorphisms appear to have a similar magnitude of association with D2 receptor BP in healthy participants.The intronic O-methyltransferase (COMT) is an enzyme that degrades dopamine, so genetic variation in the COMT gene has been hypothesized to affect dopamine receptor availability. The effect of the COMT missense variant rs4680  on striatal D2 receptor BP was examined in three independent molecular imaging studies. One study of 45 cigarette smokers found no significant effect, but reported only a lower bound on significance (P>0.25).12 A second study of 45 healthy participants also reported no significant effect ).19 A third study of 15 individuals with 22q11 deletion syndrome reported lower binding associated with the Met158 allele ).20 Thus, the available evidence does not support an effect of this polymorphism on D2 receptor BP in diploid individuals, but the genetic effects may be stronger among individuals with 22q11 deletion syndrome, who have only one copy of COMT.Catechol-SLC6A3). No significant differences between genotype groups were found among healthy controls13 or cigarette smokers.12 Among patients with traumatic brain injury, 10-repeat homozygotes showed marginally higher D2 receptor binding than 9-repeat carriers (P=0.052).13Two independent molecular imaging studies examined a 3′ variable-number tandem repeat polymorphism in the dopamine transporter gene  striatal D2 receptor BP were reported for 4 of the 15 variants. Three of the 4 variants reported were in the 560 ref. , the syn560 ref.  and the 11C] raclopride, which is displaceable by endogenous dopamine, in combination with a behavioral challenge intended to induce striatal dopamine release . Significant associations were reported for polymorphisms in genes directly involved in dopamine signaling, viz., the dopamine transporter,12 D2 receptor,26 D3 receptor27 and D4 receptor.12 In addition, significant associations were found for variants in genes for the serotonin 2C receptor28 and mu-opioid receptor37—both known to regulate dopamine release via their expression in ventral tegmental interneurons—and for variants in the genes for leptin29 and oxytocin,31 which act at their respective receptors within the mesoaccumbal and nigrostriatal pathways.Several genetic association studies used the radiotracer [DRD2 and 10 other genes have been tested for effects on in vivo striatal D2 receptor BP, but only 4 variants  have been examined in two or more comparable, independent samples. Second, despite apparent discrepancies in the literature, the ANKK1 Taq1A polymorphism (rs1800497) is robustly associated with D2 receptor BP in healthy humans. Third, the DRD2 variant rs1079597 has effects on D2 receptor BP that are similar to rs1800497. Fourth, several examples were identified in which the presence of a neuropsychiatric syndrome modified the effects of genetic variants on D2 receptor BP.This systematic review and meta-analysis has produced several principal findings. First, variants in in vivo receptor binding measures. The PET and SPECT studies examined here estimated a quantity proportional to Bmax/Kd, where Bmax is the total number of available receptor sites and 1/Kd is the affinity of the radiotracer for the receptor. Consequently, group differences may arise from a difference in Bmax, a difference in Kd or both. In addition, because raclopride is displaceable by endogenous dopamine in vivo, Bmax reflects the number of receptor sites not occupied by endogenous ligand rather than the total number of receptor sites.38 In other words, BP is sensitive to endogenous dopaminergic tone, which may be an additional source of variance. Furthermore, raclopride is known to bind D2 receptors in both the G-protein-coupled (‘high' or active) state and the non-coupled (‘low' or inactive) state, so genetic influences on the proportion of receptors in the high versus low state cannot be distinguished.39 It is notable that different tracers or modeling methods used across different laboratories will yield different absolute BP estimates . BP estimates may also be altered by between-study differences in how the striatum was defined anatomically, and by striatal atrophy in disease groups. We believe that such specific methodological factors are unlikely to substantially influence our analyses of genetic effect sizes because such factors are likely to influence genotype groups equally.Our study has notable limitations. The first type of limitation is related to interpretation of in vivo D2 receptor binding. If protocols for sharing of molecular imaging data are developed, future mega-analysis40 of individual participant data could address these and other interesting questions. Finally, relevant data may have been unreported due to negative-results reporting bias  and our meta-analysis may have overlooked relevant articles.The second type of limitation is related to systematic review and meta-analysis methodology. With the exception of rs1800497, the paucity of replications prevented reliable quantification of effect size for most genetic variants. In several cases, studies could not be included in meta-analyses because effect sizes (or data sufficient to compute effect sizes) were not reported. We used heterogeneity tests, funnel plots and sensitivity analyses to evaluate for bias and inconsistency across studies, but the utility of these approaches is limited when examining only five studies. Meta-analysis methodology is incapable of answering questions about (or controlling for) potentially interesting individual-level variables such as sex, age, race/ancestry, body mass index or striatal anatomy. Therefore, we were unable to determine with confidence whether these variables moderated the effect of rs1800497 on ANKK1 Taq1A variant (rs1800497) on striatal BP in healthy participants. BP was lower among carriers of the minor allele (‘A1' or Lys713) relative to A2 homozygotes, with a SMD of −0.57 and 95% CI of . Furthermore, the findings from in vivo molecular imaging studies were consistent with effect sizes for Bmax values measured in two postmortem studies. It is instructive to compare this effect size of 0.57 with other, commonly encountered, individual differences. Human males are typically taller and heavier than females, but the sexes overlap substantially. Among US adults, the SMD in height between men and women is ~1.2 and that for weight is ~0.4,41 so we can conclude that the Taq1A polymorphism is more strongly associated with D2 receptor BP than sex is with weight, but less so than sex is with height. It is also instructive to consider the proportion of the observed variance in BP that is explained by the Taq1A variant, as there are presumably numerous genetic and environmental factors that influence D2 receptor BP. For a SMD d with group sizes n1 and n2, the equivalent Pearson correlation is r=d/(d2+c)1/2, where c=(n1+n2)2/n1n2 . With an effect size of this magnitude, it is not surprising that some molecular imaging studies did not detect an effect of the Taq1A variant on D2 receptor BP. The largest study reviewed here included just 56 healthy participants. This sample size has only 50–60% power to detect a true effect size of 0.57 , whereas a sample of ~200 subjects provides >95% power.42Our most robust finding is the effect of the 1n2 ref. . In the 43 and a study of ~500 patients linked the A1 allele with poorer cognitive outcomes after traumatic brain injury.44 A recent meta-analysis suggested an association of the A1 allele with attention deficit hyperactivity disorder, although the authors observed significant unexplained heterogeneity across studies,45 possibly related to differences in diagnostic procedures or clinical subtypes. A recent study of ~1300 adolescents showed that A1-allele carriers had poorer performance on visuospatial working memory tasks.46 That study also measured striatal activation using functional magnetic resonance imaging during a reward task and found no significant main effects of the Taq1A polymorphism, but the authors did detect an interaction whereby working memory performance was associated with striatal activation only among A1 carriers.46 This finding is consistent with the idea that lower D2 receptor expression among A1 carriers increases the dependence of working memory performance on striatum-based motivational circuitry.47 Taken together, these studies suggest that lower expression of D2 receptors among A1 carriers confers poorer cognitive function, although this model is yet to be directly tested.Several large studies have reported associations of the A1 allele of rs1800497 with poorer cognitive function. A study of ~2000 older adults linked the A1 allele with lower general cognitive abilityANKK1 (Glu713Lys) and lies within 10 kb of DRD2.33 To our knowledge, no in vitro mechanistic studies have yet examined the effects of rs1800497 on the expression or molecular function of DRD2 or ANKK1 proteins. ANKK1 and DRD2 mRNA are both expressed in human striatum ,48 and expression of ANKK1 in rodents appears to be regulated by dopamine.50 To our knowledge, direct regulation of D2 receptors by ANKK1 has not been shown. Although a causal link between rs1800497 and D2 receptor density or affinity appears plausible, it is equally plausible that another polymorphism in LD with rs1800497 is the causal variant. For example, rs1800497 is in LD with the ANKK1 missense SNP rs7118900, which is predicted to create a new ANKK1 phosphorylation site and which has been associated with differential subcellular ANKK1 expression patterns in vitro.51 It is also in LD with the intronic DRD2 SNP rs1079597 (described above) and with the intronic DRD2 SNPs rs2283265 and rs1076560, which were reported to alter DRD2 RNA splicing.52 Given the clear association of rs1800497 with D2 receptor BP in humans that we have demonstrated here, and the multiple potentially causal SNPs in LD with rs1800497, further in vitro and in vivo mechanistic studies are warranted.Although the association of rs1800497 with neurocognitive function has been well studied, remarkably little is known about the impact of this variant at the molecular, cellular or circuit level. This SNP is predicted to cause a missense mutation in 44 but with better cognition among a more severe TBI sample.53 Along the same lines, rs1800497 was a significant predictor of cognitive performance among females with TBI, but not among males with TBI.54 Taken together with our finding that the effect of rs1800497 on D2 receptor BP is stronger in studies with a greater proportion of females, these findings suggest that this polymorphism may be more penetrant in females. Although the neurobiological basis of this sex difference is unclear, sex steroids may have a role, given evidence in rodents that estrogens and testosterone influence D2 receptor expression.57Our analyses suggest that disease status and sex may moderate the genetic effect of rs1800497 on D2 receptor BP. The robust association of the minor allele with lower D2 receptor BP that we observed among healthy subjects was not present in samples with schizophrenia, depression or traumatic brain injury (TBI) . Similarin vivo molecular imaging studies are warranted to identify the most influential variants. Third, genetic variation, sex and disease may interact to influence D2 receptor BP in complex ways, so the design and analysis of future studies that examine any one of these variables should consider effects of the other two as well. Such effects could be especially important for sexually dimorphic diseases . Our findings also raise the possibility of interactions with other variables known to influence D2 receptor BP . Ultimately, we expect that a fuller description of genetic and environmental factors that influence D2 receptor levels and related neurocognitive functions will improve our understanding of neuropsychiatric disease risk, pharmacological response and clinically relevant outcomes.Our findings carry implications for future studies of the D2 receptor. First, based on the robust evidence that the ANKK1 polymorphism rs1800497 is associated with D2 receptor binding in humans, future studies are warranted to determine whether rs1800497 is the causal variant (versus another polymorphism in LD), and to elucidate the molecular and cellular basis of this association. Second, our findings highlight the possibility that other variants of similar effect size exist. Adequately powered human


However, further investigation on large population and different ethnicities are warranted.Genetic polymorphisms in mTOR gene may be associated with cancer risk and clinical outcomes of cancer patients by affecting mTOR gene expression or its activation. However, inconsistent results have been reported. The aim of this study is to systematically evaluate the association between mTOR polymorphisms  and cancer risk as well as clinical outcome by a meta-analysis. We identified 10 eligible studies and extracted data by two investigators. Based on dominant and recessive models, odds ratio (ORs) and 95% confidence intervals (CIs) were calculated by using Stata, version 11 to evaluate the association strength. Our meta-analysis results showed that the wild genotype TT of rs2295080 polymorphism was associated with increased cancer risk under dominant model  in Chinese but not with clinical outcome parameters, while the TT genotype of rs11121704 was associated with poor clinical outcome parameters , such as death, metastasis and resistance to chemotherapy. However, rs2536 may not influence cancer susceptibility. In conclusion, this meta-analysis indicated the common polymorphisms in  Mammalian target of rapamycin (mTOR), also known as FRAP (FKBP112-rapamycin-associated protein), was originally discovered about 15 years ago in the study on the mechanism of action of rapamycin mTOR gene is located in chromosome 1q36.2, and there are 3434 genetic polymorphisms within this gene. A few polymorphisms could exert some effects by modulating transcriptional activity, miRNA binding, or splicing Recently, a number of case-control studies reported that the polymorphisms in mTOR gene were associated with individual's susceptibility to cancer risk and clinical outcome We searched the electronic database Medline to identify relevant reports by using terms “mTOR”, “polymorphism”, and “cancer” . The search was limited to English language articles. Additional studies were identified by reviewing the references of original studies. The studies included in our meta-analysis had to meet the following inclusion criteria: (1) evaluated the association of target mTOR polymorphisms and cancer risk and/or clinical outcomes in patients with cancer; (2) used case-control study or cohort study; (3) provided sufficient information for calculation of odds ratio (ORs) with 95% confidence interval (CI). The following data were extracted from each study: the first author's last name, year of publication, country of origin, type of cancers, number of genotyped cases and controls, number of cases and controls with each genotype, source of control groups  for cancer risk assessment, and prognosis parameters for clinical outcome assessment. For studies which investigated more than one clinical parameter, such as survival and response to chemotherapy, data were extracted separately for each parameter whenever possible.For control group of each study, the genotype frequency was assessed for Hardy–Weinberg equilibrium using the Chi-square test (P>0.05). We evaluated the association between the mTOR polymorphisms and cancer risk by calculating the pooled odds ratios (ORs) with 95% confidence intervals (CIs). We estimated the risks of mTOR polymorphisms on cancer by assuming dominant and recessive effects of the rear allele, respectively. Due to the limited data available, we only calculated the pooled OR under the dominant model.2-based Q-test, if the P value is greater than 0.05 of the Q-test, which indicates a lack of heterogeneity among studies, the summary OR or HR estimate of each study was calculated by the fixed-effects model Potential heterogeneity was checked by the χSensitivity analyses were performed by removing one study each time to reflect the influence of individual study on the pooled ORs.Egger's and Begg-Matzumdar tests were used to assess publication bias If the publication bias tests indicated bias existed, the Duval and Tweedie “trim and fill” method was used to adjust the bias All statistical analyses were done with Stata, version 11 .Through the primary literature research in Pubmed, 61 studies were identified for cancer risk and/or clinical outcome assessment for mTOR polymorphisms. However, after manually screening the titles and abstracts, 43 studies were excluded. The remaining 18 articles were reviewed and, 8 of them were removed due to lack of sufficient data or examing other mTOR polymorphisms but not rs2295080, rs2536 (T>C) and rs11121704 Based on genotyping data available, we noticed that there was a wide variation in the T allele frequency of mTOR rs2295080 polymorphism among cancer patients between Caucasians and Asians (Chinese and Korean),ranging from 0.311 to 0.808. Asians had the higher T allele frequency (0.777–0.808) than Caucasians (0.311–0.353).Overall, our meta-analysis results showed that the wild genotype TT of rs2295080 polymorphism was associated with increased cancer risk under dominant model   but not No significant heterogeneity was observed for all analyses except for rs2536 under recessive model (p = 0.017) . SensitiBegg's funnel plot and Egger's test were performed to evaluate the publication bias of literatures. The shapes of the funnel plots did not reveal any evidence of obvious asymmetry except for the association of rs2295080 and rs2536 with cancer risk under the recessive model, and the Egger's test also suggested that there was slight publication bias for the latter (p = 0.045) . In addiThis meta-analysis examined the association between the common genetic polymorphisms and cancer risks as well as clinical outcomes. A total of 5798 cancer patients and 6244 controls were included for cancer risk assessment and 1928 cancer patients were included for clinical outcome assessment. We found that the wild genotype TT of rs2295080 polymorphism were associated with increased cancer risk and rs11121704 TT genotype was associated with poor clinical outcomes, such as death, metastasis, resistant to chemotherapy, and toxicity. No significant association was found between rs2536 and cancer risk.http://app3.titan.uio.no/biotools/tool.php?app=snpexp) to evaluate the possible biological influence on mTOR gene expression. We found that the individuals with TT genotype had higher mTOR gene expression levels than those individuals with TC and CC genotypes, although not reaching statistical significance (p = 0.059). However, the rs11121704 polymorphism is located in intron, and it is unlikely that the rs11121704 polymorphism exert its effect by modulating mTOR gene expression, thus, additional explanation for this correlation may be due to linkage disequilibrium with other functional polymorphisms. This hypothesis is needed to be tested in future mechanistic studies.Since one group studied for the first time the germline genetic polymorphisms in the PI3K-AKT-mTOR and cancer risk as well as clinical outcomes Some limitations of this meta-analysis should be addressed. For the cancer risk assessment, all these studies included in our meta-analysis were conducted in Chinese population, and 5 studies conducted in USA were excluded due to insufficient genotyping data or not examine rs2295080 or rs2536 polymorphisms In conclusion,this meta-analysis showed that the mTOR polymorphisms (rs2295080 and rs11121704) were associated with cancer risk and clinical outcomes of cancer patients, respectively, and no any association was found for the rs2536 polymorphism. As all studies included in our meta-analysis for the assessment on cancer risk are limited in Chinese population, even for the evaluation on the clinical outcomes, only four studies conducted in China and USA were included, thus, further studies including a wider spectrum of subjects should be conducted in Caucasians and other ethnicities, which could result in comprehensive understanding of mTOR polymorphisms on cancer risk and the clinical outcomes.Checklist S1PRISMA checklist.(DOC)Click here for additional data file.


Background. We aimed to evaluate the association between maternal smoking and the occurrence of childhood refractive error and amblyopia. Methods. Relevant articles were identified from PubMed and EMBASE up to May 2015. Combined odds ratio (OR) corresponding with its 95% confidence interval (CI) was calculated to evaluate the influence of maternal smoking on childhood refractive error and amblyopia. The heterogeneity was evaluated with the Chi-square-based Q statistic and the I2 test. Potential publication bias was finally examined by Egger's test. Results. A total of 9 articles were included in this meta-analysis. The pooled OR showed that there was no significant association between maternal smoking and childhood refractive error. However, children whose mother smoked during pregnancy were 1.47 (95% CI: 1.12–1.93) times and 1.43 (95% CI: 1.23-1.66) times more likely to suffer from amblyopia and hyperopia, respectively, compared with children whose mother did not smoke, and the difference was significant. Significant heterogeneity was only found among studies involving the influence of maternal smoking on children's refractive error . No potential publication bias was detected by Egger's test. Conclusion. The meta-analysis suggests that maternal smoking is a risk factor for childhood hyperopia and amblyopia. Refractive error  and amblyopia are the leading causes of visual impairment worldwide, which are projected to affect at least one-third of the world's population by 2020 . The cruA few of environmental risk factors have been reported to be implicated in development of refractive error or amblyopia, such as family history, near work, and breastfeeding –6. NowadThe present meta-analysis extracted information from 9 studies with a total of 42,318 children by application of statistical techniques, hoping to draw a convincible conclusion about the effect of maternal smoking on the occurrence of childhood refractive error and amblyopia. via a systematic search through PubMed (http://www.ncbi.nlm.nih.gov/pubmed/) and EMBASE (http://www.embase.com) up to May 2015. Search strategy was applied as follows: [(refractive error OR (nearsightedness OR myopia) OR (farsightedness or hyperopia) OR astigmatism OR amblyopia) AND (smoking OR smoke)]. Reference lists from the included articles were also scanned for more relevant studies.Relevant articles were identifiedArticles were independently selected by two investigators. Studies were included in this meta-analysis if they met the following criteria: (i) the study included children suffering from refractive error  and/or amblyopia; (ii) children with normal vision were regarded as control; (iii) the odds ratio (OR) was provided or could be calculated for accessing association of subject refractive error or amblyopia with maternal smoking status during pregnancy; (iv) the study was published in English.Studies were excluded if they were not original research but reports, reviews, letters, or comments.The following information was independently extracted by two investigators by using a devised data extraction form from the eligible studies: the first author, publication year, study location, study time, age of the children, and so forth. Results were compared and discrepancies were resolved by discussing within our research team until a consensus was reached.Q statistic (χ2) and the I2 test [P > 0.05 and I2 < 50%), the fixed effect model was chosen for meta-analysis; otherwise, the random effects model would be used [P ≤ 0.05 was considered the existence of statistical differences.The statistical analysis was performed using RevMan 3.11. The heterogeneity among the included studies was evaluated with the Chi-square-based  be used . CombineSensitivity analysis was further conducted to detect the influence of each study on the overall effect by omitting one study at a time . PotentiBased on the search strategy, 548 potentially relevant articles (317 from EMBASE and 231 from PubMed) were finally identified. Among these articles, 299 were excluded for duplicate publication. Subsequently, 269 articles were excluded after reviewing the titles and abstracts, and 30 studies were reserved for more detailed evaluation. The full texts of the remaining articles were studied and 21 were further excluded, including 14 studies without sufficient data such as the style of refractive error  and OR with its corresponding 95% CI for maternal smoking with childhood refractive error and amblyopia and 7 studies without maternal smoking. Finally, 9 articles were retrieved for the present meta-analysis , 15–19. The main characteristics of eligible studies were listed in P < 0.05; I2 = 69.9%), and then the random effect model was chosen for meta-analysis. Children whose mother smoked during pregnancy were 1.07 times more likely to have refractive error than children whose mother did not smoke, while the result was not statistically significant (95% CI: 0.80–1.43). No potential publication bias was detected by Egger's test . Sensitivity analysis showed that the pooled result was altered after omitting the study by Iyer et al. [As shown in P = 0.782; I2 = 0.00%); therefore, the fixed effect model was chosen for meta-analysis. Based on the pooled OR, children whose mother smoked during pregnancy were 1.47 times more likely to have amblyopia than those whose mother did not smoke, and the result was statistically significant (95% CI: 1.12–1.93). No potential publication bias was detected by Egger's test . Sensitivity analysis showed that the pooled result was slightly changed  but became no longer statistically significant (P = 0.06) after omitting the study by Williams et al. [As shown in To the best of our knowledge, this is the first meta-analysis to evaluate the influence of maternal smoking on the children's refractive error and amblyopia. Our aim was to draw a convincible conclusion regarding this influence. This meta-analysis involved 9 articles and included a total of 42,318 children among which 3,282 were suffering from refractive error and 318 with amblyopia. Results demonstrated that children whose mother smoked during pregnancy were 1.47 times more likely to suffer from amblyopia compared with children whose mother did not smoke. No statistical difference was found in refractive error between the two groups of children, but the occurrence of hyperopia was found to be significantly higher in children whose mother smoked during pregnancy than those whose mother did not smoke when the studies were stratified by different type of refractive error. via antagonizing to the muscarinic acetylcholine receptors which promote axial elongation of the eye [Potential factors related to the refractive error and amblyopia, as well as the underlying mechanisms which may be responsible for the failure of emmetropization, are all not well understood. Recent studies have hypothesized that nicotinic acetylcholine receptors may be essential in refractive development, and animal models suggest that drugs blocking nicotinic acetylcholine receptors are associated with the development of refractive error , 21. Nic the eye . NeverthIn our analysis, heterogeneity was discovered among individual studies involving the influence of maternal smoking on children's refractive error. The 9 articles qualified in our meta-analysis were studied in different countries, and the prevalence of women who smoke is varied. On the other hand, race or ethnicity was reported to be associated with refractive errors and astigmatism , 23. ThuWe have to acknowledge that the present meta-analysis has several limitations. First of all, a relatively small number of articles were enrolled in this meta-analysis, which indicated that the results should be concluded cautiously. Secondly, the pooled result was reversed by sensitivity analysis, suggesting that more relevant studies are needed and updated meta-analyses should be conducted. Thirdly, meta-analysis, as a retrospective research tool, is subject to methodological deficiencies, such as lack of appropriate control group and poor representative of population. Finally, our results may be confounded by social class, diet, alcohol intake, and prematurity and some confounding effects, such as socioeconomic factors, could perhaps account for our findings that parental smoking was associated with childhood amblyopia. This information, however, is limited in the eligible studies and has not been analyzed.Despite the above limitations, we concluded that maternal smoking is a risk factor for childhood hyperopia and amblyopia. This is important because it represents a modifiable factor that can be targeted with education. However, updated meta-analyses including more studies are needed to confirm our results.


Inhaled nitric oxide (iNO) is an important therapy for acute respiratory distress syndrome (ARDS), pulmonary hypertension and pediatric hypoxemic respiratory failure. Safety concerns regarding iNO and renal dysfunction have been reported; however, there are currently no systematic reviews on this issue. Our objective was to evaluate published randomized controlled trials (RCTs) to ascertain the risk of renal dysfunction associated with iNO therapy in patients with and without ARDS.A systematic review of databases was performed to identify RCTs which compared iNO with controls up to September 2014. Effect estimates for risk ratio (RR) of acute kidney injury (AKI) were pooled using a random-effects model.Ten RCTs involving 1363 participants were included. Inhaled nitric oxide significantly increased the risk of AKI compared with controls . In the stratified analysis, a high cumulative-dose of iNO significantly increased the risk of AKI , whereas medium and low cumulative-doses did not . In subgroup analysis by study population, an increased risk of AKI was observed in patients with ARDS  but not in those without .The available data show that iNO therapy may increase the risk of renal dysfunction, especially with prolonged use and in patients with ARDS. The risk in pediatric population is unknown owing to limited data. We suggest monitoring renal function during iNO therapy, and that future trials of iNO should evaluate renal safety.The online version of this article (doi:10.1186/s13054-015-0880-2) contains supplementary material, which is available to authorized users. Nitric oxide was first reported to be an endogenous vasodilator in 1987 ,3. PreviCumulative data from physiological studies and clinical trials suggest that iNO therapy has a good safety profile -10, and This systematic review was conducted using an a priori published protocol submitted to the PROSPERO website (Registration number CRD42013005731) and reported according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) criteria . No instTwo investigators (SYR and TMH) screened studies for inclusion and independently extracted data from the included studies. Any disagreements were resolved by discussion. A standardized recording form was used for data extraction.Quality assessment of these studies was done using the Cochrane Collaboration tool for assessing the risk of bias . We asseThe primary outcome of interest was AKI as defined in the individual studies regardless of the severity of injury. The secondary outcome was AKI requiring renal replacement therapy (RRT). Risk ratio (RR) was used as a measurement of association in the analysis.Stratified analysis by study population was performed to investigate whether the risk of iNO-associated renal dysfunction varied among different patient populations. To test the dose-response relationship between iNO and the risk of renal dysfunction, stratified analysis by duration and dosage of iNO therapy was also performed.Q-statistic and I2. Heterogeneity was considered low, moderate, and high according to I2 values of <25%, 25 to 50% and >50%, respectively [P-value <0.05 was considered to be statistically significant. The data were analyzed using Stata software, version 11 .Outcome measures were pooled using a random-effects model with the inverse variance method because of anticipated heterogeneity among the included studies. We analyzed data by intention-to-treat analysis and calculated the point estimates and 95% CI of the summary effect. When one arm of a study contained no events, 0.5 was added to all cells of the two-by-two table . Sensitiectively . Publicaectively . For allThrough the electronic searches and from references, 252 citations were identified. According to our predefined inclusion and exclusion criteria, 10 RCTs involving a total of 1,363 patients were included in the final analysis ,20-28. TThe characteristics of the included studies are summarized in Table Renal dysfunction in these trials was usually defined as an excess of creatinine level to a predefined level or the need for renal replacement therapy Table . Among tP = 0.02, I2 = 0%). The study by Payen [P = 0.048, I2 = 0%). The statistical heterogeneity was low among the analyses.For the primary outcome of AKI with any severity Figure , the pooby Payen  accounteP = 0.01, I2 = 0%). The effect estimate became larger after we omitted the influential study by Payen .For the endpoint of AKI requiring renal replacement therapy Figure , iNO alsDue to the presence of sparse data and imbalance of trial size, sensitivity analysis was performed to evaluate the influence of data synthesis methods on the estimate of summary effect. The effect estimate by the Peto method was similar to that obtained by the primary analysis with a random-effects model Table .Table 2P = 0.33), did not reach statistical significance.Figure P = 0.005), but not in patients without ARDS . Among the patients with ARDS, the risk difference for AKI between iNO and control groups was 0.067 , and the number needed-to-harm to cause one additional AKI was 15.The risk of AKI associated with iNO therapy varied among the different populations Table . The risP = 0.005, four studies), whereas short-term use did not . Notably, the four studies involving the prolonged use of iNO are all ARDS studies. Table P = 0.10).To test the dose-response relationship between iNO and the risk of AKI, we performed stratified analysis by duration and dosage of iNO therapy. Prolonged use of iNO (>7 days) significantly increased the risk of AKI  are highly oxidative compounds that are produced when iNO mixes with high concentrations of oxygen in the alveoli [2 have been positively associated with the dose of iNO administered [2 may lead to cytotoxic effects on renal parenchymal cells [The mechanism of iNO-related renal dysfunction remains unclear. Table  alveoli ,11. In pnistered . It has al cells . This hyOur study has limitations. First, the included number of studies and the number of cases in individual studies are small for the non-ARDS trials. This may have limited the statistical power of our subgroup analysis and meta-regression. Second, although the level of statistical heterogeneity was very low in all of our analyses, the heterogeneity on exposure (treatment duration and dosage of iNO) should not be overlooked. Third, death was a competing risk factor in our analysis. However, we were unable to adjust for this competing risk in study-level analysis. Individual data are needed to conduct such analyses.This meta-analysis updated the evidence on the renal safety of iNO therapy in ARDS and non-ARDS populations. Our results suggest that iNO therapy substantially increases the risk of renal dysfunction. It remains unclear whether this markedly increased risk is disease-specific in ARDS or related to prolonged use of iNO. Further studies are needed to clarify this issue. We suggest that renal function should be monitored during iNO therapy and that future clinical trials of iNO should evaluate renal safety.Previous studies have shown that iNO has a good safety profile and favorable effects on renal and splanchnic perfusion; however, iNO therapy has been reported to be associated with renal dysfunction.This meta-analysis updated the evidence regarding renal safety of iNO therapy. The study suggested that the risk of iNO-associated renal dysfunction differed between ARDS and non-ARDS populations. Nitric oxide inhalation may increase the risk of renal dysfunction, especially with prolonged use and in patients with ARDS.The safety outcome of renal dysfunction was not universally reported across iNO trials, especially in non-ARDS studies. We suggest monitoring renal function during iNO therapy, and that future trials of iNO should evaluate renal safety.