Glutathione S-transferases (GSTs) belong to a group of multigene detoxification enzymes, which defend cells against oxidative stress. Tannery workers are at risk of oxidative damage that is usually detoxified by GSTs. This study investigated the genotypic frequencies of GST Mu1 (GSTM1) and GST Theta1 (GSTT1) in Bangladeshi tannery workers and healthy controls followed by their status of oxidative stress and total GST activity. Of the 188 individuals, 50.0% had both GSTM1 and GSTT1 (+/+), 12.2% had GSTM1 (+/−), 31.4% had GSTT1 (−/+) alleles, and 6.4% had null genotypes (−/−) with respect to both GSTM1 and GSTT1 alleles. Among 109 healthy controls, 54.1% were double positive, 9.2% had GSTM1 allele, 32.1% had GSTT1 allele, and 4.6% had null genotypes. Out of 79 tannery workers, 44.3% were +/+, 16.8% were +/−, 30.5% were −/+, and 8.4% were −/−. Though the polymorphic genotypes or allelic variants of GSTM1 and GSTT1 were distributed among the study subjects with different frequencies, the differences between the study groups were not statistically significant. GST activity did not vary significantly between the two groups and also among different genotypes while level of lipid peroxidation was significantly higher in tannery workers compared to controls irrespective of their GST genotypes. Three alleles exist at the locus for human GSTM1, namely, GSTM1∗A, GSTM1∗B, and GSTM1∗0, the latter corresponding to an entire gene deletion [∗0/GSTM1∗0 homozygous and lack μ GST isoenzyme activity. This has been reported to be possibly associated with an increased susceptibility to different types of cancer including bronchogenic carcinoma [Glutathione transferases (GSTs) protect cells against electrophiles and products of oxidative stress by catalyzing nucleophilic attack of the tripeptide thiol, glutathione on different reactive electrophiles, which otherwise are potentially toxic and can bind to proteins and DNA to cause cellular damage or genetic mutation . These edeletion . About 5arcinoma , and mularcinoma .θ GST that catalyzes the conjugation of halomethanes in human erythrocytes [θ GST include industrial chemicals such as methyl chloride, methyl bromide, dichloromethane, ethylene oxide, and diepoxybutane, a reactive metabolite of 1,3-butadiene. Individuals with homozygous GSTT1 gene deletion (GSTT1∗0) show a functional deficiency in Theta GST activity in erythrocytes [∗0 genotype has been shown to vary between different ethnic groups ranging from 20 to 26% in a group of South Africans to 58% among a group of Chinese people [∗0 genotype to increased cancer susceptibility, but an increased prevalence of the combined GSTM1 and GSTT1 null genotypes has been found among head and neck cancer patients [Another polymorphic gene of the same family is GSTT1 which encodes for a class hrocytes . Substrahrocytes . The pree people , 11. Some people . Few reppatients . The GSTpatients .Workers in the tannery industries directly and inhabitants living around indirectly are at risk of severe health problems due to contamination of drinking water through percolation of untreated or incompletely treated effluents into the ground . BesidesDue to the presence of many polymorphisms in GST genes, considerable attention has been bestowed in determining the association of particular allelic variants with the outcome of a variety of diseases. Among all the enzymes of GST family, two major deletion polymorphisms in GSTM1 and GSTT1 genes have shown clinical significance. These deletions result in the absence of enzyme activity, specifically in individuals with null genotypes (absence of both genes due to deletions). The null genotype  has been associated with altered risk of a variety of pathologies including cancer, cardiovascular disease, respiratory diseases –20, and A total of 190 Bangladeshi people not related to each other were enrolled in this study. Among them, 81 were tannery workers and 109 were healthy control subjects. Before enrolling in the study, each individual was informed about the viewpoint of the study and, after getting their full consent, they were included in the present study and blood samples were collected. Individuals with hypertension and cardiovascular disease related complications, diabetes, chronic kidney disease, or any other major renal diseases and being hepatitis B virus surface antigen (HbsAg) positive were excluded from this study. Healthy control subjects were nontannery workers, randomly selected from the homogeneous ethnic background as of the tannery workers. During interview session, two of the tannery workers were found to be suffering from diabetes and high blood pressure. Thus, finally blood samples from 79 tannery workers were analyzed for the study. All the subjects were also interviewed according to the structured questionnaire and the data were recorded. The study was approved by the Ethical Review Board of the Faculty of Biological Sciences, University of Dhaka, Bangladesh. Five milliliters of blood was collected from each individual in EDTA containing vacutainer tubes by the help of an expert phlebotomist in the presence of the concerned physician. Genomic DNA was extracted and quantity and quality of the extracted DNA were verified according to our previous method .β-globin (268-base pair) was amplified with the primers 5′-CAACTTCATCCACGTTCACC-3′ (forward primer) and 5′-GAAGAGCCAAGGACAGTTAC-3′ (reverse primer) as an internal control of DNA sample. The PCR was performed under the following conditions: 10 minutes of denaturation at 95°C followed by 40 cycles of 1 minute at 95°C, 1 minute at 56°C, 1 minute at 72°C, and a final extension for 10 minutes at 72°C. The amplification products were separated on 2% agarose gels, stained with ethidium bromide for the analyses of genotypes, and only those PCR signals were considered in which the corresponding β-globin gene internal control was apparent.GSTM1 and GSTT1 genes were amplified by allele specific polymerase chain reaction (PCR). Primer sequences for GSTM1 were 5′-GAACTCCCTGAAAAGCTAAAGC-3′ (forward primer) and 5′-GTTGGGCTCAAATATACGGTGG-3′ (reverse primer), which produced a 219-base pair PCR product. The GSTT1 primers were 5′-TTCCTTACTGGTCCTCACATCTC-3′ (forward primer) and 5′-TCACCGGATCATGGCCAGCA-3′ (reverse primer), which produced a 480-base pair PCR product. In the same PCR reactions, Glutathione S-transferase activity was measured according to the method described by Mannervik and Guthenberg . The reaμL serum + 900 μL saline), followed by the addition of 30 μL of 50 mM butylated hydroxytoluene (BHT). The mixture was incubated for 15 minutes in a boiling water bath. It was then kept in ice bath for 15 minutes. Above steps were repeated for the standards. Then, the sample tubes were centrifuged for 10 minutes at 2000 rpm. Finally, the supernatant was collected to measure the absorbance at 535 nm.Thiobarbituric acid reactive substances (TBARS) value was determined according to the method of Yagi . Two mL t-test was performed. A p value of less than 0.05 was considered significant. The GSTM1 and GSTTI genotype frequencies were obtained by direct counting. Statistical analyses between different genotypic subgroups of tannery workers and healthy controls were performed using paired sample t-tests.The results were expressed as mean (±SD) and, to compare the differences between different variables from the control and tannery workers, independent Student's 2) of the total study subjects were 30.95 ± 6.76 years and 23.1 ± 3.21, respectively. The data showed that 54.4% of tannery workers had primary education, 30.4% had secondary education, 12.7% had no education, and only 2.53% had higher education. On the other hand, all the healthy controls were well educated. More than 30% of the total populations were smokers. Their mean systolic blood pressure and diastolic blood pressure were 122.38 ± 13.35 and 79.52 ± 7.7 mmHg, respectively. The mean age of the healthy controls was 28.0 ± 2.0 years. Their average BMI (kg/m2) was 24.3 ± 3.34. The mean systolic and diastolic blood pressure of the control subjects were 126.30 ± 15.41 and 81.73 ± 8.80 mmHg, respectively. Twenty percent of them were smokers. A total of 79 workers were included in this study among whom 72 were males and 7 were females. The mean age of tannery workers was 33.90 ± 11.50 years. Their average BMI (kg/m2) was 21.90 ± 3.09. The mean systolic and diastolic blood pressure of the tannery workers were 118.46 ± 11.29 and 77.31 ± 6.60 mmHg, respectively. Detailed anthropometric parameters have been presented in The average age and BMI  had clear chest sound while 5.1% had noisy sound in their chests. On the other hand, when healthy control subjects were interviewed no such complications were reported.β-globin gene was used as a positive marker. Sizes of the bands were 219 bp, 268 bp, and 480 bp for GSTM1, β-globin, and GSTT1, respectively.Amplified PCR products were evaluated by agarose gel electrophoresis using a 2% agarose gel and visualized under ultraviolet light after ethidium bromide staining using gel picture analyzer . n = 12) of the total individuals were null genotypes for both GSTM1 and GSTT1 genes (−/−). Among 109 healthy controls, 54.1% had both GSTM1 and GSTT1 allele, 9.2% had GSTM1 allele with no GSTT1 allele, 32.1% had GSTT1 allele with no allele for GSTM1 gene, and 4.6% individuals represented null genotype. Out of the total 79 tannery workers, 43.3% had both GSTM1 and GSTT1 allele, 16.5% had GSTM1 allele with no allele for GSTT1 gene, 30.4% had GSTT1 allele with no GSTM1 allele, and 8.9% of the total workers represented null genotypes for both GSTM1 and GSTT1 genes. Frequency analyses of the GSTM1 and GSTT1 genotypes have been presented in Out of the total study subjects, 94 individuals (50.0%) had both GSTM1 and GSTT1 alleles (+/+) while 23 individuals (12.2%) had GSTM1 allele with null GSTT1 allele (+/−) and 59 individuals (31.4%) had no GSTM1 allele with the presence of GSTT1 allele (−/+). Further, 6.4%  as shown in p > 0.05).The mean activity of GST was estimated to be 1.73 ± 0.35 U/L in the plasma of healthy controls, while this value in the plasma of tannery workers was measured to be 1.88 ± 0.59 U/L. Statistical analysis using independent sample Out of the total study subjects, the mean activity of GST in the individuals with both GSTM1 and GSTT1 alleles was 1.92 ± 0.61 U/L, in individuals having GSTM1 allele with null GSTT1 allele it was 1.81 ± 0.43 U/L, in individuals having GSTT1 allele with null of GSTM1 allele it was 1.66 ± 0.4 U/L, and in individuals with both null genotypes for GSTM1 and GSTT1 genes it was 1.11 ± 0.46 U/L. Tannery workers with both alleles for GSTM1 and GSTT1 genes and double null genotypes (−/−) had higher activity of plasma GST (2.05 ± 0.71 and 1.95 ± 0.48 U/L) compared to their healthy counterparts . On the other hand, the plasma GST activities in the individuals either having GSTM1 allele with null allele for GSTT1 gene or having GSTT1 allele with null allele for GSTM1 gene were almost similar to those in the plasma of healthy controls . Statistical analyses revealed that the variations of GST activities in the plasma of tannery workers with different genotypes were not significant. Distributions of GST activities in different genotypic groups of study subjects have been presented in The mean level of TBARS in the plasma of the total tannery workers was 1.80 ± 0.82 nmol/mL while in the plasma of the healthy controls it was found to be 0.66 ± 0.63 nmol/mL. The level of TBARS in the plasma of tannery workers was significantly higher than that of the healthy controls as shown in p < 0.001) compared to the mean level of the healthy controls (0.65 ± 0.62 nmol/mL). On the other hand, the plasma level of TBARS in the working individuals having GSTM1 alleles with null alleles of GSTT1 gene was 1.59 ± 1.16 nmol/mL, while in their healthy counterparts the mean level was 0.92 ± 0.87 nmol/mL. Also, the level of TBARS in the plasma of tannery workers having only alleles for GSTM1 gene was 2.03 ± 0.8 nmol/mL compared to the level present in their healthy counterparts (0.64 ± 0.56 nmol/mL). It was found that the mean levels of plasma TBARS in these groups of heterozygous tannery workers were significantly higher than those of the control subjects. Moreover, the mean levels of TBARS in the plasma of the tannery workers and healthy controls having null alleles for both GSTM1 and GSTT1 genes were 1.68 ± 0.42 nmol/mL and 0.78 ± 0.08 nmol/mL, respectively  in their plasma  may contribute to the increased risk for malignancies as a consequence of a decreased efficiency to detoxify possible carcinogens. The occurrences of double null genotypes vary among different ethnic groups. In the present study, the frequency of double null genotype (−/−) was higher in the tannery workers than that of the healthy controls. Considering these facts, though 50% of the total population had both GSTM1 and GSTT1 alleles, rest of them are at risk of genotoxicity and oxidative stress due to the absence of xenobiotics detoxifying GSTM1 and/or GSTT1 alleles.The GST activity is a parameter for identifying tissue damage and neoplasia . This stThis study revealed that the genotypic variations with regard to GSTM1 and GSTT1 genes were distributed among the study subjects with different frequencies and there were no significant differences in various genotypes of GSTM1 and GSTT1 gene polymorphisms between tannery workers and healthy controls. GST activity varied among different genotypes while the rate of lipid peroxidation in tannery workers was significantly higher than that of healthy controls irrespective of their different genotypes and this indicated that the tannery workers are at high risk of oxidative damage. Moreover, grouping of tannery workers in different risk groups based on the GST activity (in addition to GSTM1 and GSTT1 genotypes) might enable us to develop a model of regular health screening for the individuals working in this toxic environment. However, more studies with higher number of participants are warranted to establish the frequency distribution of genes for other GST isoenzymes. In summary, only GSTM1 and GSTT1 alleles may not be the sole contributor to GST activity thereby instigating the importance of future studies in this area.


Previous studies have evidenced an association between gastroesophageal reflux and esophageal adenocarcinoma (EA). It is unknown to what extent these associations vary by population, age, sex, body mass index, and cigarette smoking, or whether duration and frequency of symptoms interact in predicting risk. The Barrett’s and Esophageal Adenocarcinoma Consortium (BEACON) allowed an in-depth assessment of these issues.Detailed information on heartburn and regurgitation symptoms and covariates were available from five BEACON case-control studies of EA and esophagogastric junction adenocarcinoma (EGJA). We conducted single-study multivariable logistic regressions followed by random-effects meta-analysis. Stratified analyses, meta-regressions, and sensitivity analyses were also conducted.Five studies provided 1,128 EA cases, 1,229 EGJA cases, and 4,057 controls for analysis. All summary estimates indicated positive, significant associations between heartburn/regurgitation symptoms and EA. Increasing heartburn duration was associated with increasing EA risk; odds ratios were 2.80, 3.85, and 6.24 for symptom durations of <10 years, 10 to <20 years, and ≥20 years. Associations with EGJA were slighter weaker, but still statistically significant for those with the highest exposure. Both frequency and duration of heartburn/regurgitation symptoms were independently associated with higher risk. We observed similar strengths of associations when stratified by age, sex, cigarette smoking, and body mass index.This analysis indicates that the association between heartburn/regurgitation symptoms and EA is strong, increases with increased duration and/or frequency, and is consistent across major risk factors. Weaker associations for EGJA suggest that this cancer site has a dissimilar pathogenesis or represents a mixed population of patients. National Cancer Institute. It is composed of investigators from around the world and brings together population-based case-control and cohort studies of Barrett’s esophagus, EA and EGJA. The primary objectives of BEACON are to facilitate well-powered, combined investigations of risk factors in relation to these diseases, as well as aid in the development of new studies of etiology, prevention and survival.Australian Cancer Study  FINBAR (Factors INfluencing the Barrett’s/Adenocarcinoma Relationship) study, based in Ireland Twelve BEACON studies included in a pooled analysis of tobacco smoking in relation to adenocarcinomas of the esophagus have been described previously In combination, these five studies provided 1,197 EA cases, 1,317 EGJA cases, and 4,711 population-based controls. We restricted the analytic population to white non-Hispanics, due to the relatively small number of non-White, non-Hispanic case patients . After these exclusions there remained 1,128 EA cases, 1,229 EGJA cases, and 4,057 controls for analysis. Data acquisition and data pooling for each study were approved by the Institutional Review Board or Research Ethics Committee of the institute(s) sponsoring each study.Self-reported questionnaires were administered at or near the time of cancer diagnosis for case patients and at time of recruitment for control subjects. The two primary exposures for the study were symptoms of heartburn and regurgitation. Heartburn symptoms related to burning or aching pain behind the breastbone/sternum not due to heart problems, and regurgitation symptoms were commonly specified as a sour taste resulting from regurgitation of acid, bile or other stomach contents into the mouth. The questions used by each study to ascertain these exposures are shown in Table 1 in 2]), cigarette smoking, alcohol consumption, use of proton pump inhibitors, H2 receptor antagonists, antacids, any anti-reflux medications , and non-steroidal anti-inflammatory drugs, as well as study-specific variables (study center for US Multi-center Study and country of birth for Los Angeles County Multi-ethnic Case–control Study).Other variables included in our analyses were age, sex, education, BMI  and 95% confidence intervals (CIs) of the association between exposure and outcome in each study. Second, the study-specific ORs were pooled using random-effects meta-analysis to generate summary ORs Study-specific, minimally adjusted logistic regression models included the covariates age , sex, and study-specific variables (where appropriate). In each study, we assessed whether any of the following variables changed pooled or study-specific dichotomous exposure estimates (ORs) by >10%: BMI, height, recent weight, cigarette smoking, alcohol consumption, non-steroidal anti-inflammatory drugs, education, fruit consumption, and vegetable consumption. Only pack-years of cigarette smoking altered estimates of a single study (FINBAR) by >10%. However, in addition to those variables included in the minimally adjusted models, we included the following covariates in all study-specific maximally adjusted models given previous evidence of associations between these exposures and adenocarcinomas of the esophagus: BMI  2 receptor antagonists, antacids, and any anti-reflux medications) and mid-year of recruitment using the STATA metareg command with 5,000 Monte Carlo permutations to generate each p value 2I value and its 95% uncertainty interval were used to estimate the percentage of total variation across studies due to heterogeneity 2I statistic of 0% indicates no observed heterogeneity, whereas larger values indicate increasing heterogeneity. All analyses were performed using STATA software, version 12 . All statistical tests were two-sided. P values less than 0.05 were considered to be statistically significant.To investigate potential effect-modification (and between-study heterogeneity) we conducted analyses of recurrent heartburn and/or recurrent regurgitation stratified analyses by age , sex , cigarette smoking (ever/never), and BMI  as these are known risk factors for esophageal adenocarcinomas. The statistical significance of potential effect-modifiers was assessed by a two-step analysis of product-terms using dichotomous  or continuous  variables combined with the primary exposures of interest. To further investigate between-study heterogeneity, we also conducted meta-regressions of anti-gastroesophageal reflux medications  associated with each summary risk estimate, with 2sI ranging from 47% to 84%.Associations between increasing duration and frequency of gastroesophageal reflux in relation to adenocarcinomas of the esophagus are shown in recurrent heartburn = 4.04, 95%CI: 3.13, 5.22, 2I = 39%; ORduration category 2 = 2.45, 95%CI: 1.31, 4.60, 2I = 71%; ORduration category 3 = 3.78, 95%CI: 2.75, 5.19, 2I = 0%; ORduration category 4 = 4.75, 95%CI: 3.18, 7.09, 2I = 54%; ORfrequency category 2 = 0.91, 95%CI: 0.68, 1.21, 2I = 0%; ORfrequency category 3 = 2.54, 95%CI: 1.83, 3.53, 2I = 10%; ORfrequency category 4 = 3.70, 95%CI: 2.30, 5.96, 2I = 66%; ORfrequency category 5 = 5.70, 95%CI: 4.23, 7.67, 2I = 0%). Exclusion of this study from the dichotomous recurrent/not recurrent regurgitation analysis in relation to EA also caused a reduction in estimated heterogeneity , although its exclusion had minimal impact on the moderate-to-high heterogeneity detected in the analyses of regurgitation duration and regurgitation frequency (data not shown). In analyses of recurrent heartburn and recurrent regurgitation exposures combined, the US Multi-center Study was the predominant source of the heterogeneity–although with exclusion of this study, heterogeneity for a majority of heartburn/regurgitation results remained at levels considered moderate-to-high (data not shown) and there was no effect on estimates of heartburn and regurgitation frequency. Sensitivity analyses of EGJA did not indicate any predominant source of heterogeneity.Sensitivity analyses demonstrated that the Swedish study was a major contributor to the heterogeneity in analyses of heartburn in relation to EA; excluding this study attenuated associations slightly and also lowered the heterogeneity  in relation to the association between recurrent heartburn and esophageal adenocarcinoma. Relationships for EA and EGJA were slightly stronger for women compared with equivalent estimates for men . Associations between increased frequency/duration of heartburn/regurgitation with EGJA were weaker, but still statistically significant for the highest exposed categories. From joint effects analyses, it was apparent that both increased frequency and duration of symptoms were associated with higher risk of EA. Again, equivalent analyses for EGJA exhibited similar, albeit weaker, associations.halsbränna refers to a burning sensation which could occur retrosternally and/or in the upper throat. When recurrent heartburn and recurrent regurgitation variables were combined, the US Multi-center Study was the predominant contributor of heterogeneity, although even after exclusion of this study, heterogeneity remained moderate-to-high for most summary estimates. It is conceivable that GERD may vary in its carcinogenic potency in different populations for reasons such as genetic background  and diet. For example, the composition of refluxate can affect symptom perception as well as the capacity for mucosal damage Although statistically significant associations for recurrent GERD (heartburn or regurgitation) exposure and cancer were observed separately for each study, there was moderate-to-high heterogeneity in the magnitude of the observed relative risk estimates, with the strongest associations often provided by the Swedish study. This was particularly evident for analyses of recurrent heartburn in relation to EA. High heterogeneity was also observed in a recent meta-analysis of gastroesophageal reflux and EA Associations of heartburn/regurgitation in relation to EGJA were positive, but not as strong as those observed for EA. A possible reason for this is that EGJA tumors likely represent a heterogeneous groups of malignancies–some with a pathogenesis similar to that of EA and others with a pathogenesis similar to that of gastric cancer Strengths of this analysis include the availability of individual participant data which enabled harmonization of variables and statistical models, as well as permitting flexibility of analysis. This reduces the likelihood that the heterogeneity detected was a result of differences in inclusion of covariates, modeling of covariates, or choice of statistical parameters. The consortial approach enabled generation of the largest dataset yet to permit assessment of the association between gastroesophageal reflux and adenocarcinomas of the esophagus. Limitations of this analysis include the moderate-to-high heterogeneity associated with a majority of summary estimates presented–cautious interpretation as to the magnitude of these estimates is therefore warranted. It is important to note that this pooled analysis assesses self-reported symptoms of heartburn and regurgitation, yet exposure may not always elicit symptoms. However, it has been shown that symptoms are indicative of greater severity of acid reflux exposure In conclusion, our analysis of individual participant data from the international BEACON consortium provides evidence for a strong relationship between gastroesophageal reflux exposures and adenocarcinomas of the esophagus, and indicates that longer duration and increased frequency of reflux are both associated with carcinogenic risk. Future studies should aim to ascertain gastroesophageal reflux exposures across the life-course using validated exposure assessment tools. In addition, studies are needed to further elucidate the morphological, functional, molecular and bacteriological mechanisms that link severe gastroesophageal reflux disease to cancer.File S1Study and exposure ascertainment information, and effect-modification and interaction analyses of gastroesophageal reflux in relation to adenocarcinomas of the esophagus.(DOC)Click here for additional data file.


This meta-analysis aims to examine the association between being overweight/obese and risk of meningiomas and gliomas as well as overall brain/central nervous system (CNS) tumors.Potentially eligible publications were sought in PubMed up to June 30, 2014. Random-effects meta-analysis and dose-response meta-regression analysis was conducted. Cochran Q statistic, I-squared and tau-squared were used for the assessment of between-study heterogeneity. The analysis was performed using Stata/SE version 13 statistical software.2) females seemed particularly aggravated in terms of brain/CNS tumor  and meningioma risk . In males, overweight/obesity status correlated with increased meningioma risk , whereas the respective association with overall brain/CNS tumor or glioma risk was not statistically significant. Dose-response meta-regression analysis further validated the findings.A total of 22 studies were eligible, namely 14 cohort studies  and eight case-control studies . In females, overweight status/obesity was associated with increased risk for overall brain/CNS tumors , meningiomas  and gliomas . Obese (BMI>30 kg/mOur findings highlight obesity as a risk factor for overall brain/CNS tumors, meningiomas and gliomas among females, as well as for meningiomas among males. Glioma and meningioma are the two most common primary central nervous system (CNS) tumors, representing 70% and 20% of brain tumors, respectively , 2. GlioObesity is a well established risk factor for several cancer types ; it has To our knowledge, however, no effort has been undertaken till now, to quantitatively synthesize all published cohort and case-control studies, in order to evaluate the potential association between glioma, or overall brain/CNS tumors and obesity.In view of the former considerations, our aim was to comprehensively examine the association between obesity and risk for brain/CNS tumors, glioma as well as meningioma in adults, synthesizing all available evidence. Separate analyses were conducted in males and females, in order to evaluate potential sex-specific discrepancies.This meta-analysis was performed in line with the PRISMA guidelines. Eligible studies were identified in PubMed; no restriction regarding publication language was adopted and the end-of-search date was June 30, 2014. The details about the search algorithm are provided in Cohort and case-control studies that examined the association between BMI and risk of overall brain/CNS tumors, glioma or meninigioma were deemed eligible. A systematic search in the reference lists of eligible articles and relevant reviews for potentially eligible articles was performed . In case of overlapping study populations the larger study was ultimately included. Two reviewers  independently performed the selection of studies; in case of disagreement, the final decision was reached by team consensus.The data extraction from eligible studies included: first author’s name, publication year, journal where the study was published, type of study (cohort or case-control), study period including the follow-up, geographical region, age of subjects, adjustment factors in multivariable analyses. Specifically for case-control studies, the number, definition and eligibility criteria of cases and controls, as well as the matching factors were extracted. Regarding cohort studies, the cohort size and number of incident cases were abstracted. Two reviewers  independently extracted the data and, in case of disagreement, final decision was reached by team consensus.Regarding effect estimates, the maximally adjusted Odds Ratios (ORs) for case-control studies as well as the maximally adjusted Relative Risks (RRs) for cohort studies were abstracted, together with their Confidence Intervals (CIs) for each BMI category. Separate abstraction was undertaken for males and females. When the aforementioned information was not available, the data of 2x2 tables in the articles were used for the calculation of crude effect estimates and 95% CIs.Letters were sent to the authors of studies which did not separately report relative risks pertaining to overweight and obese subjects, as well as to articles reporting exclusively on overall brain/CNS cancer, requesting the subgroup analyses on gliomas and meningiomas. The corresponding authors were contacted twice (a reminder e-mail was sent one week after the first e-mail).2 and obesity was defined as BMI >30 kg/m2. In this analysis, the term “study arms” refers to the various BMI categories presented in the included studies. The overweight and/or obese “study arms” were pooled, as appropriate. Apart from the overall pooling of overweight/obese subjects , subgroup analyses were performed for overweight and obese subjects. Separate models were constructed regarding overall brain/CNS tumors, meningiomas and gliomas, stratified by gender . The pooled effect estimate was calculated on the basis of random effects models . Cochran Q statistic and I2 were used for the assessment of between-study heterogeneity [2 as an index [2 (tau^2) were also provided.Being overweight was defined as BMI 25–30 kg/mogeneity ; neverthan index , 25 the a to b the arithmetic mean was adopted. Regarding the upper, open ended categories (≥a), two alternative approaches were undertaken [an+ (an—an-1) was implemented  and an-1 pertains to the immediately lower category (an-1to an). During pooling of study arms, the values according to the approach by Berlin et al. [n-1 category is open, as is the case in studies adopting a binary categorization.Dose-response meta-regression analysis serves the investigation of the relationship between different BMI levels and the outcomes. A specific BMI value was allocated to each study arm according to the algorithm described by our previous meta-analysis . Specifidertaken , namely:dertaken ) and , where n et al.  were pren et al. . As descn et al. , the lat2 [At the meta-regression analysis, the log of the study effect estimates was set as the dependent variable in a general linear model. BMI was analyzed in increments of 5 kg/m2 ; we pres2  and Il’y2 ) were pra priori set at 5 years regarding the desirable length of follow-up, whereas the cut-off value for completeness of follow-up was set at 85%. The studies were rated by two independent reviewers (IN-S and IGT); in case of disagreements, final decision was reached by team consensus.The Newcastle-Ottawa Quality scale  was usedConcerning the assessment of publication bias, the overall (overweight and obese pooled together) pooling analyses in males and females were chosen, in order to maximize the power of the relevant tests . The EggTaken as a whole, 14 cohort studies , 30–42 i2 = 12.8%, 2 = 24.4%). The positive association was mainly due to the obese female subjects, as increased brain/CNS tumor risk was observed among them , the pooled analysis did not yield any statistically significant association regarding brain/CNS tumor risk  was hampered by the smaller number of eligible study arms; no significant associations were noted therein  as well as case-control studies. Once again, the positive association could be mainly attributed to the obese female subjects, as increased meningioma risk was noted among them  as well as case-control studies. Statistical significance was achieved regarding overweight females . Moreover, overweight status or obesity also emerged as a risk factor for meningioma in males, conferring an excess risk of 58%.From a methodological point of view, dose-response meta-regression analysis was often hampered by the small number of eligible study arms; nevertheless, it was capable of further validating the association between being overweight/obese and increased risk of meningioma among females. Indeed, the dose-response pattern may reinforce the causal nature of the association, according to the Bradford-Hill criteria . Apart fObesity may act through a variety of cancer-promoting pathways, such as chronic insulin resistance, hyperinsulinemia and increased activity of insulin-like growth factor (IGF)-I . ImportaThe underlying mechanisms regarding the sex-specific pattern regarding gliomas and overall brain/CNS tumors remain unclear, but may encompass sex hormone-related effects. The well established, positive correlation between body fat mass and the levels of estrogen , highligLimitations of our meta-analysis essentially reflect the shortcomings of individual studies. Unadjusted effect estimates have been occasionally calculated, suffering essentially from confounding, whereas gender-specific effect estimates were not always provided, as discussed earlier. Furthermore, the number of eligible study arms regarding males was substantially smaller than that regarding females, potentially limiting the power of the analyses. Moreover, heterogeneity is very often present but may remain undetected, especially for small meta-analyses , as was It would also be tempting to envisage future studies assessing more elaborate indices of obesity, such as the waist to hip ratio as a marker of central, obesity, or measurements of subcutaneous fat, and whole body fat proportion in an attempt to provide further insight for the underlying pathophysiological links. Future studies may also broaden the perspective of this meta-analysis, examining the associations of obesity with schwannomas and other CNS tumor types, regarding which there is currently limited data. Individual patient data (IPD) pooled analyses seem also particularly desirable, as they might effectively investigate mediating and moderating effects regarding both patient and study characteristics, whereas they might effectively account for missing data as well as minimize or more adequately explain heterogeneity.Despite the limitations, this meta-analysis bears certain strengths that can also be acknowledged. The lack of publication bias might be one of them; nevertheless, tests for publication bias are rather underpowered, especially in the context of less than 10 studies . Other aIn conclusion, this meta-analysis is the first to highlight obesity as a risk factor for overall brain/CNS tumors, meningiomas and gliomas among females, as well as for meningiomas among males. Further accumulation of data seems mandatory regarding more elaborate obesity indices. These results open novel perspectives in the prevention of CNS tumors and warrant further investigation regarding the mechanisms underlying the pathophysiological links and potential sex-specific effects.S1 Checklist(DOC)Click here for additional data file.S1 Dataset(XLSX)Click here for additional data file.S1 File(DOCX)Click here for additional data file.


It is unclear whether liver transplantation is associated with an increased incidence of post-transplant head and neck cancer. This comprehensive meta-analysis evaluated the association between liver transplantation and the risk of head and neck cancer using data from all available studies.PubMed and Web of Science were systematically searched to identify all relevant publications up to March 2014. Standardized incidence ratio (SIR) and 95% confidence intervals (CIs) for risk of head and neck cancer in liver transplant recipients were calculated. Tests for heterogeneity, sensitivity, and publishing bias were also performed.Of the 964 identified articles, 10 were deemed eligible. These studies included data on 56,507 patients with a total follow-up of 129,448.9 patient-years. SIR for head and neck cancer was 3.836-fold higher  in liver transplant recipients than in the general population. No heterogeneity or publication bias was observed. Sensitivity analysis indicated that omission of any of the studies resulted in an SIR for head and neck cancer between 3.488 (95% CI: 2.379–4.598) and 4.306 (95% CI: 3.020–5.592).Liver transplant recipients are at higher risk of developing head and neck cancer than the general population. Long-teantation . The mosantation –8.Despite the rarity of head and neck cancers in the general population , approxiA systematic, comprehensive literature search was carried out using the PubMed and Web of Science databases, to identify all articles investigating the risk of head and neck cancer in liver transplant recipients . Search keywords, both free text and medical subject headings (MeSH), included ‘liver transplantation’, ‘organ transplantation’, and ‘head and neck cancer’. The search was limited to studies published in English and to those including only human subjects. Abstracts and unpublished reports were not considered.Studies were selected for meta-analysis if: (a) they were population-based cohort studies in liver transplant recipients; (b) SIR with 95% confidence intervals (CI) were calculated in transplant recipients relative to the general population; (c) they included sufficient data on the incidence, SIR or relative risk (RR) of head and neck cancer, or sufficient numbers of patients; (d) and if they assessed patients with head and neck cancer , defined as cancer of the lips, oral cavity, larynx, pharynx, nose and ear. Studies were excluded if: (a) they were case-control studies, case series, or case reports; (b) they lacked sufficient data for meta-analysis; (c) they assessed head and neck cancer following transplantation of organs other than the liver; or (d) they assessed cancers other than head and neck cancer following liver transplantation.The data from all qualified and adaptive publications were independently extracted by two investigators (QL and LY) using the inclusion criteria mentioned above; any discrepancies were resolved by discussion. Characteristics extracted from each study included the first author’s family name, year of publication, type of transplant, data source, the country of participants, number of patients, number of liver transplant cases, number of all cancers, length of follow-up time, mean follow-up time (years), patient-years (years), mean age at transplantation (years), median time to development of any type of cancer (months), number of expected cases of head and neck cancer, number of identified cases of head and neck cancer, and the SIRs of commonly known cancers and head and neck cancer.All statistical analyses were performed using STATA software . The unadjusted RR, along with the 95% CI, in each study was employed to assess the strength of head and neck cancer risks in liver transplant recipients relative to the general population.P value. A P value for the Q-statistic greater than 0.05 indicated a lack of heterogeneity among studies, allowing the use of a fixed-effects model (the Mantel–Haenszel method). Otherwise, a random-effects model (the DerSimonian and Laird method) was applied. I2 was defined as the proportion of total variation resulting from heterogeneity among studies, as opposed to random error or chance, with I2< 25%, 25–75% and >75% representing low, moderate and high degrees of inconsistency, respectively [Heterogeneity among studies, defined as differences in study outcomes, were assessed using a chi-squared-based Q-statistics test. The models of analysis for the pooled RRs were based on the ectively , 18.One-way sensitivity analysis was performed to assess the effects of individual study data on the pooled RR. Each included study was removed from the pool and the remaining studies re-analyzed to estimate the stability of the results. Begg’s funnel plot and Egger’s test were used to analyze publication bias. Funnel plot asymmetry was assessed using Egger’s linear regression method on the natural logarithm scale of the RR. A symmetric plot suggested publication bias, with a P value <0.05 indicating significant publication bias.The initial search identified 964 related articles. Application of the inclusion and exclusion criteria resulted in the retrieval of 10 eligible publications. These 10 articles, which included 13,853 liver transplants in 56,507 patients, evaluated whether the total SIR for head and neck cancer was higher in liver transplant recipients than in the general population –29. An oTable I2 = 26.2%, Pheterogeneity =0.203) in the pooled analysis.Calculation of the overall SIR of the 10 articles in our meta-analysis showed that the risk of head and neck cancer was significantly higher in liver transplant recipients than in the general population   Most of the studies were from western countries, with only one involving patients from an Asian country (Japan). Despite an exhaustive literature search and the absence of publication bias, as shown by the funnel plot and Egger’s test, some non-English publications and unpublished data related to our study may have been omitted, especially those reporting negative findings that could potentially influence the results. It was not possible to evaluate the effect of ethnicity (genetic background) on the association between head and neck cancer and liver transplantation. 2) The liver transplant recipients were not all screened for head and neck cancer before transplantation. Therefore, we could not exclude the possibility that some of these cancers may have been present prior to transplantation. 3) Because our screened studies had patients of different backgrounds, and their methods and sample size varied, some data were missing. Two indices, mean age at diagnosis of malignancy and mean time to develop head and neck cancer, were not measured in the included studies. 4) Detailed information on the type and dosage of immunosuppressive drugs was not available for further subgroup analysis. Low-dose cyclosporine was associated with a lower rate of malignancy than high-dose cyclosporine in renal transplant recipients . MoreoveOur meta-analysis showed that liver transplant recipients were at a higher risk of developing head and neck cancer than the general population. Liver transplant recipients should therefore be screened intensively for head and neck cancer, which may enhance early diagnosis and treatment, as well as patient outcomes.


INSR) polymorphisms in PCOS predisposition has attracted much attention. We performed a meta-analysis to investigate the association between the single nucleotide polymorphisms (SNPs) of INSR and PCOS. Published literature from Pubmed, Embase, and Cochrane CENTRAL was retrieved up until 7 August 2014. A total of 20 case-control studies including 23,845 controls and 17,460 PCOS cases with an average Newcastle-Ottawa quality assessment scale (NOS) score of 6.75 were analyzed. Ninety-eight SNPs distributed in 23 exons and the flanking regions of INSR were investigated, among which 17 SNPs were found to be associated with PCOS. Three SNPs detected in more than three studies were selected for further analyses. Twelve studies including 1158 controls and 1264 PCOS cases entered the analysis of rs1799817, but no significant association was found for every genotype (p > 0.05). Further subgroup stratification by ethnicity and weight did not lead to discovery of significant correlation (p > 0.05). For rs2059806, four studies including 442 controls and 524 PCOS cases were qualified for meta-analysis, and no significant association with PCOS was found for any genotype (p > 0.05). Four studies including 12,830 controls and 11,683 PCOS cases investigated the correlation between rs2059807 and PCOS, and five of the six cohorts indicated a significant impact. Our current meta-analysis suggests no significant correlation between rs1799817/rs2059806 SNPs and susceptibility of PCOS, while rs2059807 could be a promising candidate SNP that might be involved in the susceptibility of PCOS.Polycystic ovary syndrome (PCOS) is the most common gynecological endocrine disorder. The genetic background is believed to play a crucial role in the pathogenesis of PCOS. In recent years, the role of insulin receptor ( FBN3), fat and obesity associated gene (FTO), insulin (INS), insulin receptor (INSR), insulin receptor substrate 1 (ISR1), DENN/MADD domain containing 1A (DENND1A), thyroid adenoma associated protein (THADA), and luteinizing hormone receptor (LHR), may contribute to the development of PCOS [DENND1A may alter the aminopeptidase activity of endoplasmic reticulum and is involved in the development of PCOS [Polycystic ovary syndrome (PCOS) is the most common gynecological endocrine disorder that is characterized by ovarian dysfunction, hyperandrogenism, and polycystic ovary morphology . It is a of PCOS ,8. For e of PCOS .INSR gene is located at the short arm of chromosome 19 and comprises 22 exons. The Insr knockout mice manifested extreme insulin resistance [INSR and insulin resistance [INSR may impose a genetic predisposition for the development of PCOS. Indeed, a series of studies have been conducted to investigate the relationship between single nucleotide polymorphisms (SNPs) of INSR and PCOS. However, data from these studies appeared to be highly controversial [INSR and PCOS [INSR and PCOS. These findings ignited a renewed interest in the topic. Since then more studies have been performed, and both positive and negative results have been reported.sistance . Insulinsistance . Considesistance , allelicoversial ,13. Althand PCOS , two Genand PCOS  and 2012and PCOS  pointed INSR polymorphisms in PCOS, and the results often contradict each other, we carried out an updated meta-analysis covering recent data. The findings will provide useful information for elucidating the relationship between the polymorphisms of INSR and the risk of PCOS.As there have been plenty of studies exploring the role of A total of 150 articles were recognized by database searching and reference reading , among wINSR gene. We picked the SNPs that were examined in more than three studies for further meta-analysis. The PCOS patients and controls were genotyped for rs1799817 SNP in 12 studies, for rs2059806 in four studies, and for rs2059807 in four studies.The initially included studies consist of 10 from Asia, seven from Europe, two from North America, and one from South America. These studies cover totally 17,460 cases and 23,845 controls . All of As shown in As shown in  vs. CT, CC vs. TT, CT vs. TT, CT vs. CC + TT, and TT vs. CC + CT, no significant association was found, with mild between-study heterogeneity. For CC vs. CT + TT, a random effect model was selected due to severe between-study heterogeneity (p = 0.06), and no significant association was found (p = 0.41).For rs1799817 , a totalp = 0.62) while the heterogeneity from Asian population was still severe (p = 0.002). Both group indicated no significant association . Subgroup analyses stratified by weight showed that the heterogeneity in obese population was mild (p = 0.78) while in lean population was still severe (p = 0.004). The random effect model was selected, and no significant correlation was found . Only three studies carried sufficient data to be analyzed in the weight subgroups, and the data was limited to CC vs. CT + TT comparison. No correlation was found between rs1799817 and PCOS for CC vs. CT + TT, in both lean and obese subgroups. Further analyses indicated no significant difference of BMI between patients with different genotypes (p = 0.15). In addition, considering the divergence of diagnostic criteria, we performed the subgroup analysis according to the three diagnostic criteria. Consistent results were obtained from the analyses on groups diagnosed by the same criteria.To explore the origin of between-study heterogeneity, subgroup analyses with stratification by ethnicity and weight were carried out . The het vs. GA, GG vs. AA, GA vs. AA, GG vs. GA + AA, GA vs. GG + AA, and AA vs. GG + GA  and p = 0.04. No significant association was found from comparison of other genotypes, including GG vs. GA, GG vs. AA, GA vs. AA, GA vs. GG + AA, and AA vs. GG + GA, in the sensitivity analyses.Sensitivity analysis was performed by removing one study at one time. For rs1799817, removing any study in the analysis of CC  removed , a signiFunnel plots were applied to evaluate the potential publication bias. Visual inspection of the funnel plots indicated no significant asymmetry in analyses for both rs1799817  and rs20INSR gene have been implicated in PCOS, but the results were divergent. For example, for rs2252673, a study by Lee et al. [p = 0.5688) while the study by Goodarzi et al. [p = 0.006). One reason for the discrepancy could be the limited sample sizes and random errors. In this situation, a meta-analysis pooling together all the available data could effectively reduce the bias and achieve a more reliable conclusion.PCOS is associated with diversified genetic and environmental factors. Multiple candidate genes have been reported to increase the risk PCOS, but often the claims could not be confirmed in validation trials. In the past twenty years, a great number of studies were performed, and a dozen SNPs in different regions of e et al.  indicateINSR polymorphisms related to PCOS, rs1799817 SNP is the most thoroughly investigated. INSR contains α and β subunits that are encoded by exons 1–11 and exons 12–22, respectively [et al. [Among the ectively . Exons 1ectively . SNP rs1 [et al.  have perConsidering that ethnic group possess diversified genetic and environmental backgrounds, and gene-disease associations may vary across these groups , we carrINSR gene was investigated in three Caucasian cohorts and one Asian cohort. In the present study, we pooled the results from previous studies and found no evidence that rs2039806 represents a predisposition of PCOS. The sensitivity analysis for GG vs. GA + AA showed that, when a study from Iran was removed, an association with PCOS arose with a marginal p value. It should be pointed out that the positive results are only based on three studies. Further investigation is needed to clarify the impact of this specific genotype on the risk of PCOS.SNP rs2059806 located in the exon 8 of INSR gene, we noticed that five SNPs involved in the susceptibility of PCOS concentrate in exon 9 and 5' intron of exon 9, and rs2059807 is one of these SNP located in the “hot spot”. A haplotype constructed with four SNPs in this region was demonstrated to be associated with PCOS [While many significant SNPs distribute in different exons and flanking regions of ith PCOS , indicatThis study has several limitations. First, as mentioned above, since we failed to connect with some authors to collect the original data, the power of the subgroup analysis of BMI was compromised, and the pooled OR of rs2059807 could not be calculated; Second, the ethnicity of some countries could not be clearly defined and we had to classify them as the majority. Similarly, cutoff values for lean and obese were somewhat different among different studies, and we accepted the various definitions by individual studies; Third, original studies used various control groups, including healthy women, infertile women, and elderly women, and various diagnostic criteria of PCOS, making it difficult to control the confounding factors.INSR polymorphisms for genetic predisposition of PCOS. Our results suggested the absence of significant correlation between rs1799817 or rs2059806 SNPs with the development of PCOS, even if ethnicity and BMI were taken into account. However, rs2059807 could be a promising candidate SNP that might be involved in the development of PCOS. Further investigation is required to clarify the role of rs2029807. Since metabolism abnormality is a major contributing factor for the pathogenesis of PCOS, subgrouped analyses according to BMI and/or insulin resistance status may much enhance the study power.In conclusion, our meta-analyses summarized the available data concerning the role of This meta-analysis was developed according to the PRISM statement. A protocol was registered in PROSPERO, with registration number: CRD42014013145.Two authors independently searched the relevant studies published in all languages via 3 databases, including Pubmed, Embase, and Cochrane CENTRAL no later than 7 August 2014. The discrepancies were resolved by discussion. The following keywords were used: (PCOS OR “polycystic ovary syndrome” OR “polycystic ovarian syndrome”) AND (INSR OR CD220 OR HHF5 OR “insulin receptor”) AND . Additionally, we hand-searched the cited references to obtain more relevant studies.INSR SNPs and the risk of PCOS; (2) they were case-control or cohort studies; (3) sufficient data were provided to calculate odds ratio (OR) and 95% confidence interval (CI). Studies were excluded if: (1) they were family-association or case series studies; (2) the studies were from the same research group and the studied populations had overlaps. In this case, only the study with largest sample size was included for analysis and the others were excluded. Studies were included in this analysis if: (1) they evaluated the association between p values; and (12) numbers of genotypes in cases and controls. We contacted investigators for additional information when extra information was required. When the SNPs were reported with different nomenclatures, we mapped them on the chromosome and unified them into reference SNP ID numbers.The following data was extracted from every study by two reviewers independently, and the discrepancies were resolved by discussion: (1) name of the first author; (2) year of publication; (3) country; (4) ethnicity; (5) age; (6) BMI; (7) diagnosis criteria; (8) methods of genotype; (9) SNP detected; (10) OR; (11) The NOS of case control studies was applied to evaluate the quality of studies. Each study was evaluated independently by two authors , and the INSR SNPs with PCOS were assessed by calculating the ORs and mean difference among the pooled data, and the statistical significance was calculated with Z test. Q test was employed to assess the between-study heterogeneity. A p value higher than 0.1 was considered to be of no serious heterogeneity, and a fixed-effect model (FEM) was subsequently applied to calculate the parameters of the data pool [p < 0.05 was considered to be statistically significant.HWE was tested by the chi-square method. Review Manager 5.2  was used for the meta-analysis. The associations ofata pool . If the


TLR3) gene.There are inconsistent data on the association of risk of hepatitis virus infection and hepatitis virus-related diseases with the toll-like receptor 3 . Subgroup analyses by a polymorphism site indicated an increased risk of HCV infection in relation to the TT/CT genotypes of rs3775291 (1.50 [1.11–2.01]), and a decreased risk ascribed to the T allele (0.20 [0.16–0.25]). We also noted an association between rs3775291 and significantly increased risk of HBV-related diseases (2.23 [1.55–3.21]). No significant inter-study heterogeneity or publication bias was detected in the analyses.The TLR3 gene. Large-scale studies with racially diverse populations are required to validate these findings.These data suggest a likely effect on the risk to infect HCV and develop HBV-related diseases for the  A long list of conventional risk factors for HBV and HCV acquisition have previously been identified.3 In addition, genetic variations, typically single-nucleotide polymorphisms (SNP), in innate immunity receptors might be susceptibility factors for hepatitis virus infection.4Hepatitis B virus (HBV) and hepatitis C virus (HCV) are 2 ever-increasing health problems around the globe.5 Responses of the innate immune system are inducible and can be activated by pattern recognition receptors, which stimulate conserved host defense signaling pathways that regulate the expression of many immune response genes.6The innate immune system is a very important mechanism in defense against pathogenic intrusion. Delayed recognition and detection of the presence of infecting pathogens lead to organ dysfunction, inappropriate systemic responses, devastating tissue damage, life-threatening infections, and even death.TLR3 gene triggers innate immune responses, enhances the production of cytokines essential for activation of innate immunity, and recognizes pathogen-associated molecular patterns expressed on infectious micro-organisms.Toll-like receptor 3 (TLR3) is a key pattern recognition receptor of the innate immune system. The 7 Earlier research suggests significantly higher expression levels in chronic HCV infection patients with TLR3 genotypes compared to their healthy counterparts.7 Recent research also demonstrates evidence of a higher infection rate in individuals who harbor SNPs of the TLR3 gene that controls efficacy of innate immunity.8 Therefore, SNPs in TLR3 may increase the risk of hepatitis virus infection and hepatitis virus-related diseases.SNPs occur in >1% of the general population. They could induce amino acid conversions and thereby modify the promoter activities.16 But the results are inconsistent. The objective of this study was to examine the associations between TLR3 and risk of HBV infection, HCV infection, and HBV-related diseases by means of meta-analysis.Multiple SNPs positioned in the TLR3 gene, especially those under investigation , have been targeted to assess the risk of HBV-, HCV infection, and related diseases.TLR3, CD283, hepatitis, HBV infection, HCV infection, polymorphism, polymorphisms, variants, and variations. These words were used in combination or in isolation. Two of the authors singled out all publications addressing the association of interest and examined their reference lists. They also checked the references cited in related review articles to obtain additional data. The study was approved by the institutional review board. All patients provided written informed consent.We performed a systematic literature search in PubMed, Scopus, and Embase databases, without language restrictions. Search terms included toll-like receptor 3, Selection of studies eligible for inclusion was based on the following criteria defined before the commencement of publication search. Whenever 2 or more studies were overlapped in terms of subjects, we retained the most complete and informative study with a larger sample size.Included patients with HBV-infection, HCV-infection or HBV-related disease, and non-infected or healthy controls;TLR3 gene with risk of HBV infection, HCV infection, or HBV-related disease;Examined the association of SNPs in the Clearly reported the genotype frequencies of each polymorphism.Genotype distribution in controls must be in accordance with Hardy–Weinberg equilibrium (HWE).Editorials, reviews, comments, abstracts with insufficient data, and conference proceedings.Full-length articles with inaccessible genetic data.Only patients were included.Studies that deviated from HWE.Data for each study were separately extracted by 2 authors using a standard form. Disagreements were resolved by consensus including a third author. Information on the following characteristics was recorded: first author, year of publication, polymorphisms studied, infection type, disease type, number of genotypes, total cases and controls, location in which the study was conducted, ethnicity of study subjects, proportion of males, and study setting.TLR3 SNPs and risk of HBV-infection, HCV-infection, and HBV-related diseases. The ORs were calculated by comparing the mutated homozygote/heterozygote genotypes (22/12) with the wide-type genotype (11), then 22 with 12/11 in a recessive model, and 2 with 1 in an allele frequency model. Heterogeneity assumption was measured by the chi-square-based Q test,17 with P value <0.05 considered significant. The summary ORs with the corresponding 95% confidence intervals (95%CI) were calculated with the fixed-effects or the random-effects model. Random-effects summary ORs were calculated using the DerSimonian–Laird method when significant heterogeneity presented.18 The Mantel–Haenszel method was utilized to calculate fixed-effects summary ORs in case of the absence of heterogeneity.19 HWE was checked by the chi-square test in the controls. Publication bias was determined by funnel plot and Egger's test,20 a type of linear regression methodology widely used to measure funnel plot asymmetry. Sensitivity analyses were performed to examine the validity of meta-analysis results. Statistical analyses were done by the STATA software (package v.12.0). P < 0.05 was considered statistically significant.Pooled odds ratios (ORs) were estimated to assess the relationship between TLR3 SNPs, hepatitis virus infection or hepatitis virus-related diseases were excluded. Of the14 articles whose eligibility could not be determined by title and abstract review, 6 were discarded after full-text review. The specific reasons for study exclusion and inclusion are described in Figure 16 providing 3547 cases and 2797 controls. Asian subjects were used in most studies . In addition, 2 studies employed North Americans and 1 enrolled South Americans. Regarding the setting of study, 5 included healthy individuals, 2 used individuals with previous HBV clearance, and 1 included non-HCV-infected liver recipients  for allele frequency, Table There was no association observed between the .77 1.35–.31 for r.77 1.35–.31 for r.77 1.35–.31 for rTLR3 gene were associated with a significantly increased risk of HBV-related diseases. The pooled OR was greatest in the recessive model (1.77 [1.35–2.31]), and slightly lower in dominant model (1.30 [1.15–1.48]) and allele frequency model (1.28 [1.16–1.41]). The forest plot is shown in Figure Using the Mantel–Haenszel method, the polymorphisms in the In the subgroup analysis according to polymorphism site, there was a 1.56- to 2.23-fold increased risk in relation to rs3775291 [1.56 (1.30–1.88) for dominant; 2.23, (1.55–3.21) for recessive; 1.49 (1.29–1.71) for allele frequency, Table Sensitivity analyses by meta-analysis in the absence of 1 single study were conducted to examine the stability of the combined risk estimates. Statistical significance of the summary ORs was not substantially modified (data not shown). Therefore, our results are stable.P > 0.05). Egger's linear regression test provided data that supported the absence of publication bias in the meta-analysis (P > 0.05).Funnel plots were created by the standard error plotting against the OR for each study. No evidence of asymmetry was indicated in the funnel plots ,23 a replication intermediate of many viruses. As mutations in the neighboring region (Asp413) result in significantly reduced signaling activity of TLR-3, rs3775291 may prevent glycan moiety of Asp413 from binding to dsRNA or influence its glycosylation, which seems to be able to explain the reduced signaling activity induced by rs3775291.25 Again, reduction in signaling activity of TLR-3 causes higher incidence of infections and related diseases. Therefore, it is likely that the rs3775291 polymorphism is associated with the risk of HCV infection and HBV-related diseases. Our findings are also in concordance with a recent observation that HCV per se may induce downregulation of TLR3 expression on innate immune cells.29The association between rs3775291 polymorphism in the coding region of the TLR3 polymorphisms with hepatitis virus acquisition and hepatitis virus-related diseases. Al-Qahtani et al included 707 patients infected with HBV and 600 uninfected controls of Arabian descent, demonstrating a significant effect of genetic variations in the TLR3 gene on the outcome of HBV infection.9 The positive finding was confirmed in a replication study among more than 1400 subjects of Han Chinese descent.10 In a most recent analysis, Huang et al likewise suggested a significant association between TLR3 SNPs and HBV-induced diseases.11 Unlike the aforementioned analyses, Sa et al showed the absence of an association between TLR3 polymorphisms and susceptibility to HBV and HCV infection in 109 cases and 299 healthy controls of South American descent.16 Different sample sizes are a possible explanation for the conflicting results. Another important reason may relate to ethnicity variance. According to Abdel-Hady et al, HBV infection is most prevalent in Asia, Africa, Southern Europe, and Latin America.26 With respect to HCV-related infection and related diseases, the prevalence is relatively lower in economically developed countries including Africa, the West Pacific, Eastern Mediterranean, and South-East Asia, than in less developed countries including northern and western Europe, North America, and Australia (1–2% vs 5–10%).28 Based on these data, we infer that there may be ethnic differences in the influence of TLR3 polymorphisms.Several investigations have examined the association of TLR3 polymorphisms may be ethnicity-specific. It is possible that the genetic polymorphisms with a significant role in one ethnic group may not have equal impact on another. Hence analyses restricted to certain specific ethnic groups are necessary. Third, despite the potential confounding effect conferred by age, gender, lifestyle, and environmental risk factors, we were not allowed to evaluate these effects in our study as a result of data inadequacy.Some limitations should be noticed in interpreting the results. First, since most of the polymorphisms being investigated were evaluated in a low number of studies, we were not able to detect their effects individually, possibly resulting in overgeneralized results. Second, we observed significant effects when all published data were combined. However, the impact of TLR3 gene may be potential risk factors for HCV infection and HBV-related diseases. Further analyses incorporating a larger number of samples and ethnically different populations are warranted to determine the association between TLR3 and risk of hepatitis virus infection and hepatitis virus-related diseases.In summary, polymorphisms of the


Several studies have assessed the possible increased risk of ischemic heart diseases in migraineurs, drawing different conclusions.To define and update the issue of the association between migraine and ischemic heart disease we performed a systematic review and meta-analysis of the available observational studies.Electronic databases were systematically searched up to April 2014 for observational studies dealing with the risk of any form of ischemic heart disease in subjects with migraine.Out of 3,348 records, we identified 15 studies which were included in the meta-analysis. The pooled analysis indicated an increased risk of myocardial infarction  and of angina  in subjects with any migraine compared to non-migraineurs (Figure Based on our data indicating an association of migraine with myocardial infarction and angina and on previous data showing an association of migraine, and particularly migraine with aura, with an increased risk for stroke, migraine can be appropriately considered an overall risk factor for cardiovascular diseases.No conflict of interest.


Previous studies have reported the association of glutathione S-transferase M1 (GSTM1) deletion polymorphism with genetic susceptibility of lung cancer in Chinese population. However, the results remained controversial. The aim of this study was to clarify the association of GSTM1 deletion polymorphism with lung cancer risk in Chinese population. Systematic searches were performed through the search engines of Medline/Pubmed, Web of Science, EMBASE, CNKI and Wanfang Medical Online. The pooled effects were calculated by STATA 10.0 software package and Review Manager 5.0.24. Overall, we observed an association of GSTM1 deletion polymorphism with increased lung cancer risk in Chinese population (odds ratio (OR) = 1.46, 95% confidence interval (95%CI): 1.32–1.66 for null genotype vs. present genotype) based on 53 studies including 7,833 cases and 10,353 controls. We also observed an increased risk of GSTM1 null genotype for lung cancer in stratified analyses by source of control, smoking status and histological type. The findings suggest that GSTM1 deletion polymorphism may contribute to lung cancer risk in Chinese population. Further, well-designed studies with larger sample sizes are required to verify the results. The global incidence of lung cancer is 1,608,800 per year, with an annual mortality rate of 1,378,400. It was the most commonly diagnosed cancer as well as the leading cause of cancer death in males globally, and among females, it was the fourth most commonly diagnosed cancer and the second leading cause of cancer deathEpidemiological data have shown that environmental exposures such as tobacco smoking and asbestos are the main etiological factors in lung carcinogenesis2Glutathione S-transferases (GSTs) are a group of phase II detoxification enzymes which detoxify a broad range of compounds, including xenobiotics, pesticides, products of oxidative stress, chemotherapeutic drugs and carcinogens such as benzo(a)pyrene and other polycyclic aromatic hydrocarbons7891011121314151617181920212223242526272829303132333435363738394041424344454647484950515253545556575860Systematic searches were performed in Medline/Pubmed, Web of Science, EMBASE, Chinese National Knowledge Infrastructure (CNKI) and Wanfang Medical Online, with the following terms utilized: “lung cancer” or “lung tumor” or “lung carcinoma” or “non-small cell lung cancer” or “small cell lung cancer” and “polymorphism” and “GSTM1” and “Chinese” or “China”. All publications were updated to July 15, 2014. Additional relevant references quoted in the searched articles were also selected.Criteria of literature inclusion were (a) the subjects of literature must be Chinese; (b) the papers should evaluate the association of GSTM1 deletion polymorphism with lung cancer risk; (c) case-control studies or cohort studies; (d) studies should have sufficient data for estimating odds ratio (OR) with 95% confidence intervals (CI). The exclusion criteria were (a) studies without the number of case and control or other essential information and (b) reviews and repeated or overlapping studies. For repeated studies or overlapping studies, the publication with more information was selected when more than one article was identified for the same study population.In total, ninety eight published articles were identified with the association of GSTM1 deletion polymorphism with lung cancer risk in Chinese population. We reviewed all papers according to the criteria listed, above; forty one overlapping studies and four reviews were excluded. At last, fifty three original articles that focused on the association between GSTM1 deletion polymorphism and lung cancer risk in Chinese population were determined to be eligible to enter our study  and STATA10.0 software package . All the tests were two-sided, a A database was built in the light of the extracted information from selected articles. Some essential information was listed in P value ≥0.05. Therefore, a fixed-effects model was used to calculate the summary ORs for them. A random-effects model was used to calculate the summary ORs for the rest.The heterogeneity of GSTM1 null genotype vs. present genotype was analyzed for 53 selected studies. The results showed that GSTM1 null genotype vs. present genotype for squamous cell carcinoma, hospitalized patients-based control, smokers and nonsmokers had no heterogeneity with a Funnel plot was used to assess the publication bias, the shape of funnel plot seemed to be approximately symmetrical . ResultsThe sensitivity analysis was performed to determine the influence of the individual dataset on the summary ORs by consecutively excluding individual studies. The overall effects were not changed significantly when the study was homogenous for GSTM1 null genotype vs. present genotype among total population by removing some eligible studies, indicating that our results were statistically robust .GSTM1 gene is located on the short arm of chromosome 1 1p13.3)66697071607374.3666970716414243547476777879808182838485868788899091Considering that cigarette smoking is an evident risk factor for lung cancer, and that GSTM1 is involved in the metabolism of various carcinogens present in cigarette smoking, a subgroup analysis regarding smoking status was conducted. After being stratified by smoking status, the GSTM1 null genotype was associated with an increased risk of lung cancer in both smokers and nonsmokers.Lung cancer consists of at least three major histological subtypes: squamous cell carcinoma, adenocarcinoma and small cell carcinoma. It is well-known that the development of squamous cell carcinoma and small cell carcinoma is strongly correlated with cigarette smoking, whereas that of adenocarcinoma is less correlated compared with those two subtypes, which indicates that carcinogenic processes are different among the different subtypes of lung cancerThis meta-analysis should be interpreted within the context of its potential limitations. First, the combined ORs were based on individual unadjusted estimates, while a more precise analysis depending on adjusted factors should be performed if detailed individual data were available. Secondly, only published papers were enrolled in this study, which may cause publication bias. To address this issue, Egger's test and Begg's test were conducted at the same time. Our findings demonstrated that the likelihood of key publication bias might not be present in this meta-analysis. Thirdly, each study had different eligibility criteria for subjects and different source of controls, which should be taken into account while expounding the combined effects. When subgroup analysis was performed by source of control, we observed an association between GSTM1 deletion polymorphism and increased lung cancer risk in both healthy subjects-based control and hospitalized patients-based control.In conclusion, this comprehensive review demonstrates that GSTM1 null genotype might be a risk factor for lung cancer in the Chinese population. Large scale studies with the pooling of individual study data should be taken into consideration in the future studies to verify the results from this present meta-analysis.Conceived and designed the experiments: W.Y. and Y.H.; Performed the experiments: Y.S., S.F. and W.H.; Analyzed the data: Y.H. and L.J.; Contributed reagents/material/analysis tools: Y.S., S.F. and W.H.; Wrote the main manuscript text: W.Y. and Y.H.; Reference collection and data management: L.J. and Y.S.; Statistical analyses and paper writing: Y.H. and W.Y.; Study design: W.Y. and Y.H.; Prepared figures 1–4: W.H.; All authors reviewed the manuscript.


Meta-analysis is a statistical technique in which the results of two or more independent studies, with similar objectives, are mathematically combined in order to improve the reliability of the results. The outliers, which may exist even in random models, can affect the validity and strength of meta-analysis results.Ascaris lumbricoides infection.The current study uses "random effects variance shift model" to evaluate and correct the outliers in performing a meta-analysis study of the effect of albendazole in treating patients with Ascaris lumbricoides infection. The articles were published in a number of internationally indexed journals between 1983 to 2013. For both groups in each article, the total number of participants, the number of those with Ascaris lumbricoides infection, and the number of those recovered following the intake of albendazole were identified and recorded. The relative risk (RR) and variance were computed for each article individually. Then, using meta-analysis, the RR was computed for all the articles together. In order to detect outliers the "random effects variance shift model" and "likelihood ratio test" (LRT) were used. Adopting the bootstrap method, the accuracy rates for sampling distribution of the tests, which were used for multiple testing, were obtained and the relevant graphs were depicted. For data analysis, STATA and R software were used.The study used data from 14 clinical trials; each article was composed of two groups, a treatment group and a placebo group. These articles compared the effect of single dose intakes of 400 mg albendazole in treating two groups of patients with According to meta-analysis results, the estimate for RR was 2.91, with a 95% confidence interval of 2.6 to 3.25. According to the method used in this study, three articles  were outliers and, as such, they were detected in the graphs.We can detect and accommodate outliers in meta-analysis by using random effects variance shift model and likelihood ratio test. Ascaris lumbricoides infection. Ascaris lumbricoides is one of the most soil-transmitted helminthes (STH) in the world. It is estimated that 4.5 billion individuals are at risk of STH infection  and as many as 1.2 billion individuals might be infected with Ascarislumbricoides, with Ascaris lumbricoides, close to 800 million with Trichuris trichiura, and more than 700 million with hookworms , and a similar binary response variable for recovery versus nonrecovery. For each of the 14 articles, the total number of participants, the number of those infected with Ascaris lumbricoides as well as those recovered following the intake of albendazole (for each of the two groups), the effect size, and variance of the intervention were computed. Bearing in mind that each study was composed of both the albendazole and the placebo groups, the responses produced would follow a dichotomous variable. To compare the effect of albendazole on Ascaris lumbricoides, the cure rates were used to compare two groups under study. The priority index of the effect of albendazole as cure ratio in intervention group to the placebo group was considered the relative risk (RR). The effect size or RR is shown by Ɵ; then test statistic has to be defined for the significance of the effect size. Test statistics is defined by Q = Σ Wi (yi − Ɵˆ) 2 in which Wi = 1 / Vi and Ɵˆ = Σ (WiYi) / Σ (Wi). Under the null hypothesis, for all the effect sizes, which were similar or symmetrical, the distribution of Q statistics was chi-square with K-1 degree of freedom  where τ2 is the between-study variance, which is unknown, e represents residual errors with, e ~ N  where R = diag . The elements of R, the study variances, are regarded as known. The variance-covariance matrix of (1) can then be written as var(y) = V = τ2In + R with the variance of the i th study treatment effect given as var(yi) = τ2 + δ2i.Where ith study  takes the form:According to Gumedze and Jackson , the ranjdj to model (1), where dj is the jth unit vector of length n, i.e. with value 1 in the ith position and zero elsewhere, and is an unknown random coefficient with δj ~ N  for ω2j≥ 0.This adds an extra term δj indicates which study has an inflated variance. Model (2) has the form of a simple linear mixed model with δj as a random effect with variance ω2j. The variance-covariance matrix for the data under the RVSOM for the jth observation is:The subscript An extension of model (2), which allows different inflated variances for more than one study, can be written as:r} of studies considered to be outliers, D = [dj] is an n × r matrix containing entires of 0 and 1, where an entry of 1 in the ith row and jth column indicates that study i has the jth of r inflated variances, and δI is a r × 1 vector of unknown random effects. We referred to this model as an ‘extended RVSOM’  was used to measure the size or magnitude of ωˆ 2j.At First, we used forest plot diagram to detect outliers in our data, then we entered the outliers detected in forest plot in the RVSOM Model as the jth observation. Then the model was fitted to the data and the degree of ωˆ0:ω 2j = 0 against the alternative hypothesis was HA(j):ω2j > 0 for a RVSOM for observation j. Stram and Lee  to the data to obtain estimates μ ^ and τˆ2.Step 2. Generate a new data vector from model (1) and estimates μ ^ and τˆn, by fitting the model (2) to the simulated data and compute and save the order statistics of the set LRT j for j = 1,..., n.Step 3. Compute the likelihood ratio test statistics LRT j, j = 1,..., R times. This step generates an empirical distribution of size R for each order statistic.Step 4. Repeat steps 2 and 3 α. The percentiles, using α = 0.05 and k = 1 for largest order statistic, k = 2 for second largest order statistic, are shown in the plots given in the results.Step 5. Calculate the 100 (1-α)th percentile for each order statistic for the required significance level Ascaris lumbricoides infection; yi indicated the relative risk in the ith article. As the forest plot diagram  and our findings greatly alleviated any concerns about the potential impact of outliers in these data. Once again, our proposed methodology performed well.Ascaris lumbricoides, detected articles 4, 7, and 12 as outliers. The articles under study were published in the internationally indexed journals during 1983 to 2013. The study used RVSOM to detect outliers. The results obtained from the tests and the relevant figures might indicate that the studies in rows 4, 7, and 12 served as outliers in this meta-analytic review and that the findings were largely reliable. Of course, with application of the method used in article 30, the articles 2, 10, and 14 were also detected as outliers. In this regard, in another research, a meta-analysis study was conducted to investigate the effect of albendazole on two groups of case and control patients afflicted with Trichuris trichiura. The results indicated a relative risk estimate of 2.06 with 95% confidence intervals of 2.4 to 1.76. Therefore, with such results in mind, it can be claimed that, the cure ratio of individuals infected with Trichuris trichiura using albendazole was two times more than those who did not use the drug (Ascaris lumbricoides using albendazole was nearly three times more than those who did not take the drug. Therefore, it can be concluded that albendazole has been more effective in treating Ascaris lumbricoides than curing Trichuris trichiura.The results obtained from meta-analytic evaluation of 14 studies investigating the effect of albendazole on patients with the drug , 29, 30.


Clostridiales and Erysipelotrichales orders were more abundant among cases after controling for multiple testing (p = 0.011). No such difference was observed between mid-esophagitis and healthy controls. This study is the first to show that composition of gastric corpus mucosal microbiota differs in early ESCC and ESD from healthy esophagus.Observational studies revealed a relationship between changes in gastric mucosa and risk of esophageal squamous cell carcinoma (ESCC) which suggested a possible role for gastric microbiota in ESCC carcinogenesis. In this study we aimed to compare pattern of gastric corpus microbiota in ESCC with normal esophagus. Cases were included subjects with early ESCC (stage I–II) and esophageal squamous dysplasia (ESD) as the cancer precursor. Control groups included age and sex-matched subjects with mid-esophagus esophagitis (diseased-control), and histologically normal esophagus . DNA was extracted from snap-frozen gastric corpus tissues and 16S rRNA was sequenced on GS-FLX Titanium. After noise removal, an average of 3004 reads per sample was obtained from 93 subjects. We applied principal coordinate analysis to ordinate distances from beta diversity data. Pattern of gastric microbiota using Unifrac (p = 0.004) and weighted Unifrac distances (p = 0.018) statistically varied between cases and healthy controls. Sequences were aligned to SILVA database and  H. pyloriH. pylori and ESCC58910111213H. pylori negative stomach high abundance of Streptococcus and PrevotellaStreptococcus species are resistant to low pHCancer of the esophagus affects more than 450,000 people each year, of which 90% are squamous cell carcinomas (ESCC)3Golestan province in Northern Iran is located in “esophageal cancer belt”19Firmicutes, Bacteroidetes Proteobacteria, Actinobacteria, and Fusobacteria. Phyla composition was consistent across ESCC, ESD, healthy esophagus and mid-esophagitis groups. Clostridiales (FDR = 0.011) and Erysipelotrichales Orders (FDR = 0.011) in the case group compared to the healthy esophagus group  was lower than healthy controls (mean = 1449) (p = 0.03) and diseased-controls (mean = 1715) (p = 0.02).OTUs were assigned to 31 Phyla, 53 Classes, 90 Orders, 168 Families, and 336 Genera. Five most abundant phyla were us group . Helicobp = 0.003) based on 37 matched pairs. Similar difference was found when OTU abundances were counted in the weighted Unifrac distances (p = 0.018). Removing ESCC cases and restricting the analysis to esophageal squamous dysplasia, did not change the results (based on 17 matched pairs). There was no individual taxonomic difference between the diseased-controls and healthy controls. We did not observe a significant difference in the coordinates based on Unifrac distances between diseased-control and healthy-control groups.Unifrac and weighted Unifrac distances were ordinated by applying PCoA. We have observed significant differences in microbiota composition of gastric fundal mucosa in subjects with early ESCC and ESD compared to those with a normal esophagus.So far no single or combination of environmental or genetic risk factors has been identified to explain the high incidence of ESCC in Asian Cancer belt. Although alcohol and tobacco consumption account for major proportion of the disease in low risk area, their contribution to the risk of ESCC in high incidence regions of Asian Cancer Belt – from where 90% of cases arise- is limited. One suggested possible explanation for this excess risk is intrinsic exposure to carcinogens. Gastric mucosal changes have been shown to be associated with ESCC risk and bacterial alteration in the stomach has been considered to be the possible link. This study is the first to evaluate the relevance of microbial link in this association.Proteobacteria, Firmicutes, Bacteroidetes, Actinobacteria, and Fusobacteria. Our data suggests that presence and abundance of Clostridiales and Erysipelotrichales Orders were associated with early ESCC and ESD. Both microbial hints belong to Firmicutes phylum. In an animal study, presence of members of Clostridiales altered the pathogenicity of H. pylori by recruitment of CD4 T-cells to the gastric mucosaClostridiales might enhance more aggressive response to H. pylori and induces severe or pangastritis which may further develop to atrophic gastritis. The link between Clostridiales and the esophagus has been observed in response to proton pump inhibitor use, as abundance of its members increases in the lower esophagus. Such an effect has not been reported in gastric mucosa. Erysipelotrichales is highly dependent on the fat content in the diet of humanized mice modelSimilar to studies on gastric mucosa microbiotaH. pylori and ESCC risk is controversial. Most studies have not shown an excess risk including in our study populationH. pylori seropositivityHelicobacteriacea order among case group which might be secondary to subclinical atrophic changes in the stomach. Our finding on non-distinct microbial clusters in gastric mucosa in relation to different abundances of H. pylori among ESCC is similar to the report in healthy stomachH. pylori infection is prevalent in catchment area (seropositivity >84% among older >55 years), any inference on relation between H. pylori and gastric microbiota in this study should be done with caution.The relation between species levelgenus or species nomenclature to the OTUs. Although 16S is a gold standard for bacterial phylogeny, presence of its multiple copies in some bacteria with slight difference in sequence could lead to identifying multiple types of the same bacteriumIn this study, 16S rRNA with coverage of 464 nt of region V3–V4 was used to distinguish bacteria. Although it did not cover whole hyper variable V1–V9 regions, the most divergent regions were covered and it was possible to align with SILVA reference with 90.5% coverage. Furthermore, V4 and V5 are less specific for Some of gastric microbiota are anaerobes and difficult to cultivate and some of them could not be assigned to reference databases. We tried to reduce this effect through OTU formation. In our data unassigned OTUs were unevenly distributed among samples, as five samples shared most of them. These rare species could be formed due to intrinsic errors of pyrosequencingModest sample size is one of the limitations of this study, though based on national cancer registry data less than 9% of ESCC cases are diagnosed at early stagesSimilar to every PCR-based methods, amplification bias is a concern as some less abundant taxa might be underestimated or ignored. Using endoscopy clinic controls may limit the generalizability of our results to healthy populations, although several biopsies from the esophagus were carefully examined before assigning subjects to healthy control group.We used mid-esophagus esophagitis as a diseased-control to evaluate the specificity of our findings. Inflammation is a common pathway between neoplasia and esophagitis meanwhile esophagitis does not play a role in ESCC carcinogenesis. Detection of explicitly different microbiota pattern between cases and healthy controls and not with diseased-controls increases the validity of our findings. Number of esophagitis controls was less than healthy controls and 16 out of 17 were male, although no difference in microbiota was observed between male and female, reduced sample size might affect the precision of statistical tests.In this study – similar to the most other studies- we used one snap frozen biopsy from mid corpus of the gastric greater curvature. In an early study of gastric biotaThe different patterns of gastric microbiota that we observed might be secondary to cancer. We used early-stage ESCC and patients with ESD (the asymptomatic precursor lesion of ESCC) as cases, to decrease this possibility. As a sensitivity test we excluded Early-stage ESCC and observed different microbiota pattern between healthy controls and ESD subjects (data was not shown), which lowered the possibility that our findings is secondary to cancer. Although it is probable that microbiota pattern changes secondary to dysplasia.Clostridiales and Erysipelotrichales Orders. Studies with higher sequencing depth and larger sample sizes are warranted.In summary we observed an altered gastric corpus microbiota in patients with early ESCC or ESD, compared to subjects with healthy esophagus, with higher abundance of bacteria in the From all subjects, an informed consent was obtained. The case-control study was approved by the local ethical committee in the Digestive Disease Research Centre of Tehran University of Medical Sciences, Iran (IRB00001641), Institutional Review Board of National Cancer Institute, US and Stockholm Regional Ethics Vetting Board, Sweden (Dnr:2010/471-31/4). The methods were carried out in accordance with the approved guidelines.Details of study design have been reported earlierSample size of original case-control study was calculated based on achieving power of 95% to detect odds ratio of 2 between matched cases and control and total of 300 cases and 600 controls were calculated. For present study, the difference between 2 taxonomic levels of more than 20% or odds ratio of 2.5 would require at least 30 pair of case-control to achieve 90% power. A priori for odds ratio was based on reports from other studies in gastrointestinal malignancies and microbiotaAmong 579 endoscopy clinic controls and 300 ESCC cases, we found 19 early-stage ESCC and 18 ESD patients. One age/sex-matched control was randomly selected from endoscopy clinic controls with a healthy esophagus for each case. We also found 17 subjects with mid-esophageal esophagitis who were age and sex matched with the cases, and included them as a diseased control group. For each subject, one snap-frozen biopsy from mid corpus of the stomach greater curvature was preserved in RNA-later preserved at −70 and send to Karolinska Institutet on dry ice for microbiota assessment.DNA was extracted from snap frozen gastric tissue (approximately 5–6 mm in length) using DNeasy Blood & Tissue Kit . Tissues were treated with filtered-lysozyme in lysis buffer (Tris-EDTA-Triton) and overnight incubation at 56°C in buffer AL  and proteinase K and blended with glass beads (Tactum Lab. Sweden) with 0.1. 0.5 and 1 mm diameter for 1 minute in Bullet Blender . Mixture was incubated with RNase . Tubes containing only beads, lysozyme, lysis buffer, and extraction kit substances were included as quality controls for contamination.Small subunit of ribosomal RNA (16 S rRNA) gene was amplified from extracted DNA by using primers targeting region V3–V4. Forward primer (Bekt_341F: 5′-CCTACGGGNGGCWGCAG) and reverse primer (Bekt_805R: 5′-GACTACHVGGGTATCTAATCC) carried 454-adaptor sequences A and BStandard flowgrams were split to individual samples. AmpliconNoise v1.25 with its default filtering setting (PCR noise precision = 30. pyro noise precision = 0.6) was used to remove noisy dataH. pylori influences the microbiota composition, an average-linkage hierarchical clustering analysis was performed excluding H. pylori from microbiota and proportion of different quartiles of H. pylori was compared between clusters. R and phyloseq package were used for statistical analysis (www.r-project.org)Two sets of analysis were done:(a) Individual taxonomy comparison: Assigned OTUs were compared individually between study groups using conditional logistic regression with false discovery rate (FDR) control to correct for multiple comparisonsConception and design: W.Y., D.N., R.M. Development and Methodology: D.N. and W.Y. Acquisition of data: D.N., M.S., R.M., S.N.M., R.S., S.F. and F.I. Laboratory analysis: D.N. and W.Y. Analysis and interpretation of data: D.N., A.P. and W.Y. Drafting Manuscript: D.N. Review, revision of manuscript: W.Y., D.N., B.W., A.P., R.M., S.D., C.A., F.K., P.Bo. and P.Br. Study Supervision: W.Y. and R.M.