The risk of an individual to develop an acute kidney injury (AKI), or its severity, cannot be reliably predicted by common clinical risk factors. Whether genetic risk factors have an explanatory role poses an interesting question, however. Thus, we conducted a systematic literature review regarding genetic predisposition to AKI or outcome of AKI patients.We searched Ovid SP (MEDLINE) and EMBASE databases and found 4027 references to AKI. Based on titles and abstracts, we approved 37 articles for further analysis. Nine were published only as abstracts, leaving 28 original articles in the final analysis. We extracted the first author, year of publication, study design, clinical setting, number of studied patients, patients with AKI, ethnicity of patients, studied polymorphisms, endpoints, AKI definition, phenotype, significant findings, and data for quality scoring from each article. We summarized the findings and scored the quality of articles.The articles were quite heterogeneous and of moderate quality (mean 6.4 of 10).Despite different gene polymorphisms with suggested associations with development or severity or outcome of AKI, definitive conclusions would require replication of associations in independent cohort studies and, preferably a hypothesis-free study design.The online version of this article (doi:10.1186/s12882-015-0190-6) contains supplementary material, which is available to authorized users. Approximately 40 % of critically ill patients develop acute kidney injury (AKI) , 2. DespNo reliable method exists that is able to predict the individual risk to develop, or to survive, AKI. Genetic variability offers an intriguing research focus, since clinical risk factors explain only part of the conferred risk . Other cIndividual susceptibility to develop AKI and to influence disease progression have been studied using candidate gene association studies. The literature has been evaluated in two systematic reviews. Lu and colleagues  searchedAccordingly, we sought to update the current literature about genetic association to AKI with special emphasis on the quality of the studies.We systematically searched the OvidMedline and Embase databases for relevant articles from the year 2000 until March 5, 2015. The databases were selected according to the recommendation of The Human Genome Epidemiology Network (HuGENet) guidelines . We condAfter the removal of duplicates, the search resulted in 4027 references. Articles presenting an original study about human genetic variability and its possible association with AKI were selected for further analysis. We screened titles and abstracts and categorized the articles according to the predefined inclusion and exclusion criteria. Exclusion criteria were determined in an attempt to explore the most common contributors to AKI in adult subjects in a hospital setting. Exclusion was either because of the studied subjects  or methods used . In addition, a number of articles were excluded because they did not study AKI or genetic variability in association with AKI, but focused on chronic renal failure, diabetes, polycystic kidney disease, thrombotic thrombosytopenic purpura, haemolytic uremic syndrome, systemic lupus erythematosus, liver failure, predictive biomarkers, glomerulonephritides, viral infection, malignancies, or hereditary syndromes affecting kidneys. The process of article selection is presented in Fig. Two authors (L.V and M.K) independently reviewed 39 articles. Of the 39, 9 were published only as conference abstracts –20, and We extracted the following data from each article: The first author, year of publication, study design, clinical setting, number of all genotyped subjects, number of patients, number of patients with AKI, ethnicity, studied polymorphisms, endpoints, definition of AKI, intermediate phenotype, and significant findings. In addition, we assessed the study quality according to the scoring by Clark et al. . We gradThe 28 articles in the final analysis were published between 2000 and 2014. The characteristics of the articles are described in detail in Table The number of the patients studied varied between 61 and 1741 median 262) with a total of 12272. The number of patients with AKI was between 2 and 459 (median 115), but not stated in three studies –26. In s2 with a N = 26, 93 %) of the studies provided adequate information about the primer sequences of the genotyping assay.The mean quality score was 6.4 (range 3–10).  of the studies reported power calculations , 33–37 aThe correlation between the two reviewers before consensus was good (Spearman’s rho coefficient 0.90). The mean score for each of the ten quality criteria is presented in Fig. Of the 28 studies in our review, six (21 %) , 43 had The main results are presented in Fig. interleukin 6 (IL-6) gene −572 G/C polymorphism (rs1800796) or interleukin 10 (IL-10) gene −592 C/A polymorphism (rs1800872) and acute renal insufficiency. Stafford-Smith et al. [angiotensinogen (AGT) gene +842C –allele (rs699) and IL-6–572C –allele in Caucasians is associated with renal dysfunction.Jouan et al.  could noh et al.  found thtumor necrosis factor alpha (TNFA) gene −308 G/A polymorphism (rs1800629) was found by Jaber et al. to be associated with higher levels of TNF-α ex vivo and higher mortality in patients receiving renal replacement therapy (RRT) [IL-10 polymorphism −1082 G/A intermediate/high producer (GA + GG) genotype (rs1800896) was associated with higher levels of IL-10 and lower mortality in the same patient group [TNFA gene −308 polymorphism, as well as the G-allele of IL-10–1082 polymorphism were more prevalent among patients with contrast-induced acute kidney injury. Furthermore, the low-producing genotypes of the aforementioned polymorphisms were associated with the risk of AKI, as were the haplotypes T-C-A of TNFA and C-A-C-G of IL-10. Dalboni et al. [TNFA −308 polymorphism and AA genotype of IL-10–1082 polymorphism, when combined, are associated with AKI and/or death, and also with RRT and/or death. Susantitaphong et al. [TNFA gene −308 G/A A-allele was associated with higher peak serum creatinine and urinary kidney injury molecule-1 (KIM-1), as well as higher multi organ failure (MOF) score in hospitalized patients with AKI. The TNFA gene was the most investigated, with both positive and negative results  . They alnt group . Chang ent group  found thi et al.  showed treceptor IIa of the Fc portion of immunoglobulin G (FCGR2A-H131) allele in the rs1801274 polymorphism and development of acute renal failure in patients with pneumococcal community-associated pneumonia.In a study by Sole-Violan et al. , an assoangiotensin converting enzyme (ACE) insertion/deletion (I/D) polymorphism (rs4646994) [ACE D–allele and an increased risk of postoperative AKI after coronary artery bypass grafting (CABG). Contradictory to this, du Cheyron et al. [Six studies investigated 4646994) , 40, 42.4646994)  were abln et al.  found thn et al. , 40 or an et al. , 42.catechol-O-methyltransferase (COMT) gene. The low-activity genotype (LL) was associated with AKI and more frequent administration of furosemide and RRT. However, results by Albert et al. [COMT polymorphism and AKI. Association between the C-allele of the endothelial NO synthase (eNOS) -786 T/C polymorphism (rs2070744), low creatinine clearance, and more frequent administration of RRT has also been reported [phenylethanolamine N-methyltransferase (PNMT) gene rs5638 + 1543 G –allele is associated with an increased risk for AKI, and the genotype +1543 G/A is associated with oliguria, whereas PNMT rs876493 -161 A –allele is associated with lower mortality and less circulatory shock.Haase-Fielitz  studied t et al.  and Kornt et al.  suggest reported . Alam etreported  found thapolipoprotein E (APOE) have appeared to associate with a higher maximum creatinine [hypoxia-inducible factor-1alpha (HIF-1α) gene is associated with higher use of dialysis.The non-e4-allele of polymorphisms rs429358 and rs7412 of eatinine , 43, andeatinine , in two eatinine , 44. A sNADPH (Nicotinamide Adenosine Dinucleotide Phosphate) oxidase p22phox gene T-allele, associated with more prolonged length of hospital stay, higher dialysis use, and higher odds for composite outcome [In 2007, Perianayagam and colleagues found a +242 C/T polymorphism (rs4673) in the pro-oxidant enzyme  outcome .myeloperoxidase (MPO) gene −765 T/C (rs2243828), +9890 A/C (rs2071409), and +2149 T/C (rs2759) polymorphisms and lower urine output as well as the composite outcome of dialysis administration or in-hospital death, and with MPO +157 G/T polymorphism (rs7208693) and more advanced AKI as well as the aforementioned composite outcome.In 2012, Perianayagam et al.  found anCytochrome b245, α subunit (CYBA) gene, was investigated in a study by Perianayagam et al. [m et al.  who founerythropoietin (EPO) gene rs1617640 T/G -polymorphism TT-genotype to be associated with increased creatine phosphokinase-MB (CPK-MB) and administration of RRT, but no association with AKI was found [Popov found the as found .human leukocyte antigen-DR-beta (HLA-DRB) alleles and less administration of RRT, but not with development of AKI.Payen and colleagues  found ansuppressor of fused homolog (SUFU) gene, correlated with renal function in hospitalized patients with Enterobacteriacea sepsis. Vascular endothelial growth factor (VEGF) gene polymorphism +936 (rs3025039) CC-genotype was found associated with AKI by Cardinal-Fernández [In a study by Henao-Martínez et al.,  the polyernández .B-cell CLL/lymphoma 2 (BCL2) gene and a single polymorphism (rs2093266) in a serpin peptidase inhibitor, clade A  member 4 (SERPINA4) gene were found to protect from AKI, and a polymorphism (rs625145) in salt-inducible kinase 3 (SIK3) gene was found to be associated with increased risk for AKI.In a study by Frank , two polPolymorphisms studied for their association with the development of AKI and published only as abstracts are summarized in Additional file In this systematic review we identified 28 studies examining genetic predisposition to the development and/or outcome of AKI. Based on a previously-published quality-scoring method utilized and further developed in this review, these few identified studies had inadequate quality (mean score 6.4/10). The studies were heterogeneous in defining concepts and outcomes, replicated with ambiguous results, and underpowered. All but one of the reviewed studies used a candidate gene approach. Typically, associations found with such an approach have been difficult to replicate and the coverage of genetic variation has been poor . RegardlTNFA gene is suspected to alter the inflammatory response and to be associated with AKI susceptibility [TNFA and susceptibility or outcome of AKI. Second, polymorphisms rs429358 and rs7412 in the APOE–gene have been investigated among patients undergoing cardiac surgery. Apolipoprotein E is a protein that is known to function in lipid metabolism but has other roles as well. The APOE-e4–allele is suggested to be protective against AKI [ACE, providing a key enzyme in the renin-angiotensin-aldosterone–pathway, with an effect on renal blood flow, is a relevant candidate gene and has been investigated in several studies [Some generalizations may be made based on the study designs, however. First, the balance between pro- and anti-inflammatory mediators was among the most investigated. Polymorphism rs1800629 in the tibility , 45, as tibility , 30, 41.tibility , 42, 44 inst AKI , 38, 43,inst AKI . The invinst AKI , 54. Thi studies , 42. Unf studies , 38 and  studies , 40, 42 Within the assessed studies, the best-fulfilled quality criteria were: reporting statistics, reporting primer sequence, and defining the case group Fig. . Study sAlthough the initial consensus between the two reviewers was good in this study, the reviewers had several discussions about the specific quality criteria during the process. In general, the original criteria in the reviewed studies were rather subjective and, thus, hard to interpret. Furthermore, the drawback of using the criteria by Clark and colleagues  was thatBCL2, SIK3, and SERPINA4-SERPINA5 genes, suggesting that apoptotic pathways might have a role in determining the susceptibility to AKI.The study by Frank and colleagues  avoided The inferences based on this review are subject to some limitations. First, the literature search was conducted in the English language. However, the databases used were selected on the basis of HuGENet guidelines, with assistance from an information specialist. Second, no unpublished data was included in the search. Third, all the possible genetic analyses made by the researchers may not have been published in the articles and, thus, may have led to reporting or publication bias. This may also influence the understanding of the relationship between positive and negative results. Fourth, data proved to be inadequate to allow any meta-analysis. Thus, our findings are limited to the assessment of the quality and description of the individual studies. However, 22 79 %) and 16 (57 %) of the 28 studies in this review were not included in the two respective previous reviews  % and 16, an obviIn this systematic review we found 28 genetic studies focusing on association with development and/or severity of AKI. Of note, the majority of these studies lacked adequate quality. Conclusive evidence would still require a standard definition of AKI, unbiased outcome measures and replication of associations in a large independent sample and, preferably, applying a genome-wide hypothesis-free study design in future research. This systematic review was reported according to the principles in the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement (Additional file


Background. Several kinds of foods are hypothesized to be potential factors contributing to the variation of prostate cancer (PCa) incidence. But the effect of poultry on PCa is still inconsistent and no quantitative assessment has been published up to date. So we conducted this meta-analysis to clarify the association between them.Materials and Methods. We conducted a literature search of PubMed and Embase for studies examining the association between poultry consumption and PCa up to June, 2015. Pooled risk ratio (RR) and corresponding 95% confidence interval (CI) of the highest versus lowest poultry consumption categories were calculated by fixed-effect model or random-effect model.Results. A total of 27 (12 cohort and 15 case-control) studies comprising 23,703 cases and 469,986 noncases were eligible for inclusion. The summary RR of total PCa incidence was 1.03 (95% CI [0.95–1.11]) for the highest versus lowest categories of poultry intake. The heterogeneity between studies was not statistically significant . Synthesized analysis of 11 studies on high stage PCa and 8 studies on chicken exposure also demonstrated null association. We also did not obtain significant association in the subgroup of cohort study , as well as in the subgroups of population-based case-control study and hospital-based case-control study. Then the studies were divided into three geographic groups: Western countries, Asia and South America. The pooled RRs in these areas did not reveal statistically significant association between poultry and PCa.Conclusions. This meta-analysis suggests no association between poultry consumption and PCa risk. Further well-designed studies are warranted to confirm the result. Prostate cancer (PCa) is the second most common form of cancer and the fifth leading cause of cancer death among males. The age-standardized incidence rate varies up to 25-fold worldwide, ranging from 111.6 in Oceania to 4.5 in South-Central Asia per 100,000 . DietaryMeat attracted nutritionists’ concern for hazardous component. Researchers have found that meat cooked in high temperature contained considerable amount of mutagens. The hazardous component evoked the hypothesis that meat consumption would be in positive association with elevated PCa risk. Red meat and processed meat were proved to support the hypothesis and lower consumption was thus recommended . The rolTo our knowledge, no article that attempted to quantitatively synthesize the effect of poultry consumption on PCa risk has been published up to now. Although there were two existing reports involving this topic , both ofTo ensure the rigor of this meta-analysis, we designed and reported it adhering to the criteria set out by PRISMA statement . To idenStudies were eligible for inclusion if they satisfied the following criteria: (1) used a cohort or case-control design; (2) contained data on PCa incidence; (3) identified poultry as exposure, including chicken, turkey, processed poultry etc.; and (4) provided the extreme categories of poultry consumption and corresponding relative risks with 95% confidence intervals (CI).Articles in inappropriate formats, such as review, conference paper, editorial and letter, were not considered. Cross-sectional study, ecologic study and non-human research were excluded for unavailable effect size estimate. Studies that reported data for broad classification of meat, such as white meat, were excluded. If there was overlapping period of data collection in the same setting in several articles, only the most recent and complete one was chosen.Data were extracted by two independent researchers (Wan ZC and Xu XB), and the discrepancies between them were resolved by a third contributor. For each study meeting the inclusion criteria, the following characteristics were collected: author, year, study location, study design, sample size , study period, duration of follow-up for cohort studies, participants’ age at baseline, exposure, analytical comparison , the relative risk and 95% CI that reflected the maximum extent of adjustment for potential confounders, and variables adjusted for or matched by in the statistical analysis. If an article separately reported effect size estimates on total PCa and high stage PCa, both of them were extracted. Crude effect size was calculated by available data if it was absent in the original paper. When the full article or necessary data were unavailable, we attempted to contact the author by email to ask for sufficient information.Quality assessment of included studies was evaluated using the Newcastle-Ottawa Quality Assessment Scale . The scaStatistical analysis was based on comparison of the highest poultry intake category with the lowest one. As different studies reported different cutoffs , the categories were extracted from the corresponding exposure groups in original articles without unification. The measure of interest was risk ratio (RR) with 95% CI. Hazard ratio was directly considered as RR. Given that PCa was a rare outcome and thus odds ratio in case-control study mathematically approximated to RR , odds raχ2-based Q test and I-square (I2) statistic. Heterogeneity was considered significant if P value was below 0.05 for Q statistic or I2 value was above 75%. I2 statistic represented the proportion of variation across studies due to between-study heterogeneity, and the values of 25%, 50% and 75% were regarded as cut-off points for low, moderate and high degrees of heterogeneity (P < 0.2) was conducted to explore source of heterogeneity if it was significant. The following pre-specified study-level covariates were assessed: type of exposure , study design , geographic region , and adjustment for family history of PCa (yes and no). Tau-square (τ2) statistic would be used as the measurement of proportion of heterogeneity that covariate accounted for. In order to assess the influence of individual study on the overall estimate, sensitivity analysis was carried out by omitting one study at a time and calculating the summary effect for the remainders. Publication bias was evaluated by funnel plot, and quantified by Egger’s test .We also pooled the studies on high stage PCa and the studies on chicken exposure respectively. Stratified analyses by study design and geography were subsequently done. A two-tailed Detailed process of literature selection was described in P = 0.085, I2 = 28.5%). Pooled RRs from sensitivity analysis did not change substantially, which indicated that the current result was statistically reliable. The shape of funnel plot did not show any evidence of obvious asymmetry. And, no evidence of publication bias was revealed by Egger’s test (P = 0.768) and Begg’s test (P = 0.868). Separated groups that were restricted to studies with high stage PCa and studies with chicken exposure were synthesized to evaluate the effect on histology-specific PCa incidence and the effect of type-specific poultry. High stage cancer in the present study was comparably defined as Stage IIB or above in the American Joint Committee on Cancer TNM system, although various terminologies  may be used in original articles. Pooled result the 11 studies on high stage PCa showed no elevated risk  with no significant heterogeneity   . The het= 31.4%) . There w= 31.4%) . In sensP = 0.010, I2 = 66.9%). Then a series univariate meta-regression analyses were conducted by adding single covariate at a time, and only the model including adjustment for family history of PCa (yes vs. no) obtained significant result (P = 0.128). τ2 value indicated that this covariate accounted for 71.7% of the heterogeneity. In this subgroup, five of the six studies did not adjusted for family history, and one did . Of note, this analysis should be interpreted cautiously for limited studies in the model.The stratified analysis by study design also showed no elevated risk . All of  one did . After eP = 0.03, I2 = 41.6%). In the univariate meta-regression analyses, none of the pre-specified covariates were related to the strength of the association between poultry and PCa incidence. The Begg’s test, Egger’s test and visual inspection of the funnel plot did not suggest publication bias in Western countries subgroup.Furthermore, we evaluated the relation between poultry and total PCa in three geographic subgroups  accordinTo date, little is known about the etiology of prostate cancer except for a few non-modifiable factors: age, ethnicity/geography and family history. This systematic meta-analysis, based on maximal number of published studies, highlighted the effect of poultry consumption on PCa risk. Studies on total PCa, high stage PCa and chicken exposure were quantitatively synthesized. Stratified subgroups by restricting the analyses to studies with different study designs and in different geographic regions were also performed. Ultimately, we found no evidence of association between them.Although there were several studies claiming significant effect of poultry on PCa, they couldn’t reverse the holistic conclusion. Indeed, the majority of involving studies have found null association regardless of histological stage, exposure type, study design, and study location. Two earlier articles, a dose–response analysis  and a quWe considered several potential reasons for the lack of an overall association between poultry consumption and PCa incidence. First of all, meat mutagens, such as heterocyclic amines (HCAs) and polycyclic aromatic hydrocarbons (PAHs), were the most commonly mentioned mechanism which possibly led to the adverse outcome . ResearcWe revealed evidence of between-study heterogeneity in several subgroups. Whether the study adjusted for family history was proved to be a significant covariate in meta-regression analysis. Prostate-specific antigen (PSA) screening was likely to play similar role. Because, on the one hand, men who enrolled in intensive screening tended to take up healthy lifestyle. On the other hand, screening trial have demonstrated an increase in PCa incidence with PSA test . HoweverOur research had several limitations as with other meta-analyses. Firstly, diverse scales of poultry consumption and definitions of the highest and lowest categories were used in different studies. They acted as obstacles to perform dose–response analysis and caused unambiguous interpretation of the conclusion to some extent. Secondly, food frequency questionnaires were used in all included studies to assess dietary exposure, but some of them did not be validated. In addition, several studies with low quality scores may weaken the test of hypothesis.In summary, our comprehensive analysis of available data provides no evidence of association between poultry consumption and PCa incidence. Further well-designed studies are warranted to confirm the result.10.7717/peerj.1646/supp-1Table S1Click here for additional data file.10.7717/peerj.1646/supp-2Table S2Click here for additional data file.10.7717/peerj.1646/supp-3Supplemental Information 3Click here for additional data file.


The CCAAT/enhancer binding proteins (CEBPs) have been involved in the etiology of acute leukemia (AL) and investigated in numerous genetic association studies, however, the results were inconclusive. The current meta-analysis was conducted to clarify the effect of CEBPE rs2239633 variant on childhood AL risk. Electronic literature search was performed on August 15, 2014, from databases of Medline, PubMed, Embase, and Web of Science. A total of 22 case-control studies were eligible for the pooled analysis. The results demonstrated that rs2239633 A allele was significantly associated with a decreased risk of childhood AL , especially in B-cell ALL subgroup , but not among T-cell ALL or AML subgroups. In the stratified analysis by ethnicity, the association was observed in Europeans  but not in Asian and mixed populations. Moreover, the results of sensitivity and cumulative meta-analysis indicated the robustness of our results. Also, Begg’s and Egger’s tests did not indicate any evidence of obvious asymmetry. In summary, our study provided evidence that CEBPE rs2239633 variant is associated with decreased risk of childhood B-cell ALL in Europeans. Acute leukemia (AL), characterized by dysregulated clonal expansion of immature lymphoid or myeloid progenitor cells, is the most common cancer in children. In the UCEBPs) are transcription factors involved in hematopoietic cell development, including granulopoiesis. Akagiet al [CEBPB and CEBPE double-knockout mice were highly susceptible to fatal infections and died within 2–3 months. Also, the proportion of hematopoietic progenitor cells in the bone marrow of the knockout mice was significantly increased. Recently, genome wide association (GWA) studies have identified candidate single nucleotide polymorphism (SNP) located in CEBPE (14q11.2), which was strongly related to the susceptibility of childhood ALL.[CEBPE might appear to be a good candidate gene for childhood AL. Up to date, an accumulating number of studies focused on the association between CEBPE variant and ALL risk, however, the conclusions of these studies were inconsistent. Thus, we conducted a meta-analysis with an overall larger sample size by summarizing previous case-control studies to clarify the associations of polymorphisms in the CEBPE gene with susceptibility to childhood AL, including B-cell ALL, T-cell ALL and AML.The CCAAT/enhancer binding proteins  (the last search update was August 15 2014). Moreover, the reference lists of review articles and retrieved articles were hand-searched for additional potential studies.A comprehensive literature search of the Medline, Pubmed, Embase, and Web of Sciencerepositories was carried out using the following keywords: ,  and (‘CEBPE rs2239633 variant and childhood AL risk (2)providing sufficient data to estimate odds ratios (ORs) with 95% confidence intervals (95% CIs). A study was excluded if: (1) had no control population (2) investigated the adult acute leukemia.Studies meeting the following criteria were included in themeta-analysis: (1) reporting the association between The following information was gathered for each eligible study by two independent authors: name of the first author, year of publication, number of patients and healthy controls, sex and mean age in patients and healthy controls, country of origin, ethnicity of the individuals involved, method of genotyping, types of acute leukemia , allele frequency and genotype frequency. The corresponding author would be contacted when the article did not provide sufficient genotype distributions. Moreover, disagreements were resolved by discussion between the two investigators.CEBPE rs2239633 polymorphism and risk of childhood acute leukemia. The significance of the pooled OR was determined by the Z-test, and a P value of less than 0.05 was considered significant. In the control group, the Hardy-Weinberg equilibrium (HWE) was assessed, and a P<0.05 was considered as significant disequilibrium. In addition, subgroup analysis was carried out stratified by types of AL and ethnicity. Between-study heterogeneity was evaluated using the χ2-based Q test and I2 test.[2>50% was considered significant heterogeneity, and a random effects model (DerSimonian and Laird method) was used, otherwise a fixed effects model was applied.[Odds ratios (ORs) and corresponding 95% confidence intervals (CIs) were estimated to assess the association between  I2 test.,9 A P<0. applied. Subgroup applied.,12Sensitivity analysis by omitting one study at a time was also performed to assess the influence of individual studies on the combined risk estimate. We also CEBPE rs2239633 polymorphism and childhood AL risk. The selection process is illustrated in Eighty-four references were identified during our premature searches, of which, 63 non-relevant articles were excluded following review of title and abstract. In the remaining 21 full text papers, 10 studies did not investigate the association between CEBPE polymorphism with acute leukemia involving 6152 patients and 11739 healthy controls is shown in The summary of meta-analysis for the 2, respectively. Moreover, Galbraith plot analysis indicated studies as outliers, which were possible sources of heterogeneity, when excluded, the heterogeneity was non-significant (Ph = 0.77 for A vs G) and the association was still significant . , 52.8% (p = 0.01), 57.1% (p = 0.006), and 31.2% (p = 0.03) of τ<0.001). In the sensitivity analysis, the pooled OR did not qualitatively change by omitting a single study at a time.  In additPublication bias was assessed by Begg’s and Egger’s tests. The shapes of the funnel plots did not indicate any evidence of obvious asymmetry.  MoreoverCEBPE, along with other 4 CEBP family members, have been targeted by recurrent immunoglobulin heavy chain translocations in B-cell precursor ALL, suggesting a possible role of CEBPE insusceptibility to ALL.[CEBPE rs2239633polymorphism and childhood ALL has been investigated in several genetic association studies, though the results of these studies were contradictory and inconclusive, owing to the small sample sizes and ethnic difference.[Recent studies showed that y to ALL. Moreoverfference.,21,24,25fference.,27CEBPE variant and childhood ALL risk in the overall population. In the subgroup analysis by ethnicity, the results indicated that CEBPE rs2239633 polymorphism was significantly associated with childhood AL in Europeans, but not among mixed and Asian populations, suggesting that the relative contribution of individual susceptibility genes may vary across different populations. Previous studies demonstrated that racial and ethnic disparities persist in the development of ALL.[CEBPE variant with susceptibility to T-cell ALL or AML. Here, in the subgroup analysis, significant association was not observed in T-cell ALL  and AML  subgroups. Childhood AL is a group of diseases with varied immunophenotypes and genetic changes. It is suspected that B-cell or T-cell ALL and AML may not share a common etiology.[CEBPE polymorphism was only associated with susceptibility to B-cell ALL subtype, which was also supported by the evidence that GSTM1 and XRCC1 Arg399Glnvariants were only associated with ALL but not AML.[CEBPE polymorphisms in childhood AL, especially in T-cell ALL and AML.In the present meta-analysis, we found a significant association between t of ALL. In genert of ALL. Additiont of ALL. When thet of ALL. However,etiology. In the Uetiology. while thetiology.–37 These not AML.,39 Howev2, respectively, indicating that these factors were the potential sources of between-study heterogeneity. In addition, the Galbraith plot demonstrated five studies that were the potential origin of heterogeneitysince, when excluded, the heterogeneity was removed. In these studies, disease subtype heterogeneity [In our meta-analysis, heterogeneity was detected in most genetic models. Thus, we carried out meta-regression and Galbraith plot analyses to assess the sources contributing to the heterogeneity. The results of meta-regression showed that types of disease, source of control, ethnicity and year of publication could explain 79%, 52.8%, 57.1%, and 31.2% of τogeneity ,40 and eThis meta-analysis increased statistical power by pooling data from all eligible studies, whereas several limitations should be acknowledged in our meta-analysis. First, we only included studies published in English, which might introduce a language bias. Second, sample size was small for some subgroups, such as T-cell ALL and AML subgroups, which might limit the precision of the pooled estimates, suggesting that more large sample sizes, precise and stratified studies Click here for additional data file.


Recent epidemiological studies suggest that treatment with insulin may promote cancer growth. The present systematic review and meta-analysis of published observational studies was conducted to assess the risk of cancer during treatment with insulin.A compressive search was conducted through MEDLINE, PubMed, Web of Science, EMBASE, and Chinese Biomedical Literature databases (CBM). Pooled relative risks (RRs) and 95% confidence intervals (CIs) were calculated with a random-effects model.A total of four studies with one case-controls study and three cohort studies comparing the insulin therapy and colorectal cancer susceptibility were identified. When all four studies were analyzed, the summary RRs were 1.61 (95% CI = 1.18–1.35) in a random-effects model for individuals with insulin therapy, compared with individuals without insulin therapy, which suggests a statistically significant association between insulin use and colorectal cancer.Our findings provides the evidence that insulin therapy may contribute to the risk of colorectal cancer.http://www.diagnosticpathology.diagnomx.eu/vs/9339731010859509The virtual slide(s) for this article can be found here:  Colorectal cancer (CRC) is a leading cause of cancer-related morbidity and mortality in the Western world. Obesity, Western-style diet, and lack of physical activity are established risk factors for CRC . Few epiIn line with these considerations, it has been hypothesized that insulin use might influence CRC development. In the past years, several eligible case–control or cohort studies were performed to identify the association of insulin use with CRC risk. However, the results remain inconclusive and inconsistent. To date, no meta-analysis has been conducted to investigate the association between insulin therapy and CRC susceptibility. Hence, a meta-analysis based on a total of four independent studies was performed, which may provide the evidence for association of insulin therapy with CRC susceptibility.We conducted a comprehensive search using key words “DM”, “diabetes”, “CRC”, “colorectal neoplasm”, “colon cancer”, “colon neoplasm”, “rectal cancer”, “rectal neoplasm”, “insulin” and “insulin therapy” in the following key electronic biomedical databases MEDLINE, PubMed, Web of Science, EMBASE, and Chinese Biomedical Literature databases (CBM) without date and language restrictions, and all eligible studies were detected before September 2013. Their reference lists were hand-searched to find other relevant publications. Titles and available abstracts were scanned for relevance, identifying papers requiring further consideration. Of the studies with overlapping data published by the same investigators, only the most recent or complete study was included in this meta-analysis.To be included in the meta-analysis, a published study had to meet the following criteria: (1) only original articles of a quantitative assessment of the relationship of insulin therapy and risk of colorectal cancer, (2) cohort studies, (3) adult human population, and (4) insulin or one of insulin products as the main independent variable, (5) results expressed as relative risk (RR), (6) studies with a 95% CI for RR, or sufficient data to calculate these numbers. While for the exclusion criteria, we provided as follows: (1) case–control or case-only studies, family-based studies, case reports, editorials, and review articles ; (2) controls with other types of tumors, and (3) studies that did not provide data that allowed calculation of standard errors for effect estimates and if the estimates were not adjusted for age and gender were excluded. Moreover, when there were multiple publications from the same population or cohort, only data from the most recent report were included. In studies with overlapping cases/controls, the higher quality score, or the study with more information on origin of cases/controls was included in the meta-analysis.Information was carefully extracted from all eligible publications independently by two authors according to the inclusion criteria listed above. Disagreement was resolved by discussion between the two authors. The following data were collected from each study: the first author’s last name, year of publication, and country of population studied, study design, number of exposed and unexposed subjects, follow-up period, age, gender, type of DM (type 2 or combined type 1 and type 2), risk estimates with their corresponding confidence intervals, and variables controlled for by matching or in the multivariable model, numbers of cases and controls with the insulin therapy and CRC, respectively. For each study, we extracted the risk estimates that reflected the greatest degree of control for potential confounders. We did not limit the number of patients to include a study in our meta-analysis.I2 statistic, which takes values between 0% and 100% with higher values denoting greater degree of heterogeneity  [We used the crude ORs or RRs with their corresponding 95% CI as the metric of choice. Based on the individual ORs or ORs, the pooled OR was estimated. Summary RR estimates with their corresponding 95% CIs were derived by the method of DerSimonian and Laird  using bogeneity) . In addigeneity) ,12. SensThrough literature search and selection, a total of four studies with one case-controls study and three cohort studies ,4,13,14 In order to compare the difference and evaluate the sensitivity of the meta-analysis, we conducted one-way sensitivity analysis to evaluate the stability of the meta-analysis. The statistical significance of the results was not altered when any single study was omitted, confirming the stability of the results (data was not shown). Hence, results of the sensitivity analysis suggest that the data in this meta-analysis are relatively stable and credible after exclusion small size of studies.For publication bias assessing, Begg’s and Egger’s test were employed. Begg’s funnel plot was performed to assess the publication bias of the literature. The shapes of the funnel plots did not reveal significant evidence of obvious asymmetry. Furthermore, Egger’s test was used to provide statistical evidence for funnel plot symmetry. The results still did not suggest any evidence of publication bias.In this present work, one case–control study  and threSimilar to other systematic reviews and meta-analysis, our study also has some limitations: (1) small sample size is limitation in our work, (2) publication bias may be present, although we have used our best efforts to identify relevant reports/articles, (3) most of the studies did not distinguish between types 1 and 2 diabetes, (4) diabetes is an under-diagnosed disease, and some degree of misclassification of exposure to diabetes is likely to have occurred, and (5) meta-analysis is just a statistical test that is subject to many methodological restrictions and is not able to control for other relevant factors.Some limitations listed above, there are also some advantages should be emphasized. Our results may have important clinical and public health implications. DM is a serious and growing health problem worldwide. CRC, one of the most common malignancies of the gastrointestinal tract, is a significant health problem. Meta-analysis provided a popular method for combining world literatures across studies to resolve the statistical power and discrepancy problem in associate studies -18. The In conclusion, our analysis suggests that insulin therapy will significantly increase the risk of colorectal cancer. More well-designed with larger sample studies will be needed to verify our results.The authors declare that they have no competing interests.LW, SC, ZT, XZ, XC, and XB carried out the meta-analysis study, drafted the manuscript and involved in revising the manuscript critically for important intellectual content. LW, SC, ZT, and XB participated in the design of the study and revised the manuscript. LW, XZ, XC, and XB carried out the meta-analysis study and drafted the manuscript. LW, SC, ZT, XZ, XC, and XB participated in the design of the study, drafted the manuscript and revised the manuscript. All authors read and approved the final manuscript.


The association between maternal smoking during pregnancy and autism spectrum disorder (ASD) risk in offspring has been investigated in several studies, but the evidence is not conclusive. We, therefore, conducted this meta-analysis to explore whether an association exists between maternal smoking during pregnancy and ASD risk in offspring. We searched PubMed, Embase, Web of Science, and the Cochrane Library for studies of maternal smoking during pregnancy and ASD risk in offspring up to 10 June 2015. The random-effects model was used to combine results from individual studies. 15 observational studies (6 cohort studies and 9 case-control studies), with 17,890 ASD cases and 1,810,258 participants were included for analysis. The pooled odds ratio (OR) was 1.02 : 0.93–1.13) comparing mothers who smoked during pregnancy with those who did not. Subgroup and sensitivity analysis suggested the overall result of this analysis was robust. Results from this meta-analysis indicate that maternal smoking during pregnancy is not associated with ASD risk in offspring. Further well-designed cohort studies are needed to confirm the present findings. Autism Spectrum Disorders (ASDs) are a group of heterogeneous neurodevelopmental conditions characterized by social and communication difficulties, and by restricted and repetitive interests and behaviors . The ASDet al. concluded that there was no association between maternal prenatal smoking and ASD in offspring based on 15 observational studies [et al. [et al. misunderstood the difference between the period of “prenatal” and “pregnancy”, and they inappropriately included studies that only reported smoking during pregnancy but not specified whether it was during the prenatal period or not. We therefore performed this study to systematically assess the association of maternal smoking during pregnancy  and ASDs risk.Smoking during pregnancy may lead to numerous adverse outcomes in the offspring, such as low birth weight, placenta abruption, and sudden infant death syndrome . Materna studies . However [et al.  only seaThis meta-analysis was performed according to the Meta-Analysis of Observational Studies in Epidemiology guidelines . A systeStudies were included if they met the following criteria: (i) had a cohort or case-control design; (ii) reported the association between maternal smoking during pregnancy  and ASD risk in their offspring; and (iii) reported risk estimates and 95% confidence interval (CI). Only the study with the most complete data was included if data was reported more than once in different literatures. Non-human studies, letters, case reports, conference abstract, review, and studies that with insufficient information were excluded.Two reviewers independently extracted data, any disagreements were resolved by discussion. The following data were extracted: last name of the first author, publication year, study design, study location, age, sex, number of cases and total population, risk estimates adjusted for the maximum number of confounding variables with corresponding 95% CIs, and confounding factors that were being adjusted for in the analysis. We also extracted any dose-relationship data for smoking and ASDs risk.The methodological quality was assessed using the Newcastle-Ottawa Scale (NOS) . The NOSp < 0.10) and the I2 statistic [p > 0.10 and I2 ≤ 50%, otherwise the fixed effect model was chosen [We combined the results using the OR as a measure of the association between smoking and ASD risk. Heterogeneity was investigated by the Cochrane Q statistic . The pooled OR estimates based on 15 studies showed that maternal smoking during pregnancy was not associated with an increased risk of ASDs .The meta-analysis data that explored the effect of maternal smoking during pregnancy on the risk of ASDs is shown in et al. study [et al. study [l. study  to 1.05(l. study .p > 0.05).Begg’s test and Egger’s test suggested no significant publication bias among the included studies (all The present study used a comprehensive search strategy and explored the relationship between maternal smoking during pregnancy and ASDs based on 15 observational studies. However, the overall results did not suggest a significant association between maternal smoking during pregnancy and ASDs risk.Several hypotheses of association between maternal smoking during pregnancy and ASD are plausible. First, nicotine is the chemical that most probably has adverse effects on brain development among thousands of ingredients of tobacco smoke  and its Despite all the hypotheses mentioned above, the pooled OR failed to find a positive association between maternal smoking during pregnancy and ASDs. However, in the subgroup analysis, a significant negative association was observed in the North America. Although the association was not statistically significant in the individual studies of this subgroup, the increased sample size narrowed the confidence interval of the pooled result and thus it became statistically significant. Nevertheless, further studies are needed to examine this contrary finding, considering that the results were not fully adjusted and potential bias may exist due to the design of included studies.Our study has several limitations that deserve consideration. First, because this meta-analysis was based on observational studies, our results were prone to recall and selection bias inherent in the included studies. In addition, the potential confounding from other risk factors could not be ruled out, especially when considering that some studies only reported crude risk estimated without adjustments. Second, potential misclassification of maternal smoking may lead to incorrect estimates of associations. The smoking status was generally collected at early pregnancy and was based on self-report, there was a possibility that participants tend to narrow their smoking status and the smoking habit might change during pregnancy. Third, high statistical heterogeneity was found across the studies that may weaken the strength of our findings; however, the subgroup analysis, as well as sensitivity analysis, showed that results from our meta-analysis was robust. Finally, the included studies of our meta-analysis were mainly conducted in the North Europe and the North America. The findings of this study had limited implications for other populations, like the Africans and Asians, as both the genetic and environmental factors in relation to autism varied among different ethnicities.This meta-analysis indicates that there is no association between maternal smoking during pregnancy and ASD. Given the limited number of cohort studies included in this study, the present findings should be confirmed in further prospective cohort studies.


Studies investigating the association between hepatitis C virus (HCV) infections and the occurrence of cholangiocarcinoma (CCA), especially intrahepatic cholangiocarcinoma (ICC), have shown inconsistent findings. Although previous meta-analyses referred to HCV and CCA, they mainly focused on ICC rather than CCA or extrahepatic cholangiocarcinoma (ECC). Since then, relevant new studies have been published on the association between HCV and ICC. Since the different anatomic locations of CCA have distinct epidemiologic features and different risk factors, it is necessary to evaluate the relationship between HCV infection and ICC, ECC, and CCA.Relevant studies were identified by searching PUBMED, EMBASE, and MEDLINE databases prior to 1 August 2013. Pooled risk estimates were calculated with random-effects models using STATA 11.0.versus 2.01). The Begg funnel plot and Egger test showed no evidence of publication bias.A total of 16 case-control studies were included in the final analysis. Pooled risk estimates showed a statistically significant increasing risk of CCA (odds ratio (OR) = 5.44, 95% CI, 2.72 to 10.89). The pooled risk estimate of ICC  was higher than that of ECC . In a subgroup analysis, the pooled risk estimate of ICC in studies from North America was obviously higher than in Asia (6.48 HCV infection is associated with the increasing risk of CCA, especially ICC. Cholangiocarcinoma (CCA) is a malignant neoplasm arising from the epithelial cells of the intrahepatic or extrahepatic bile ducts. Therefore, CCA can be classified as intrahepatic cholangiocarcinoma (ICC) and extrahepatic cholangiocarcinoma (ECC). CCA is the second most common primary hepatic malignancy, representing 10% to 25% of primary hepatic malignancies worldwide, accounting for 3% of all gastrointestinal tumors .Defining the risk factors for CCA benefits the search for better ways to prevent the occurrence of this disease. To date, it has been shown that hepatolithiasis, primary sclerosing cholangitis, liver flukes, biliary duct cysts, specific toxins , inflammatory bowel disease, and genetic polymorphisms are the major risk factors for CCA -8. AlthoTwo independent reviewers searched PUBMED, EMBASE, and MEDLINE databases prior to 1 August 2013. The MeSH  search included headings that were all combinations of the hepatitis C virus such as: HCV, cholangiocarcinoma, intrahepatic, extrahepatic, bile duct cancer, and bile duct neoplasm. We also reviewed the reference lists of retrieved articles to search for more relevant studies. There were no language restrictions.Studies were eligible if they met the following criteria: (1) case-control studies; (2) exposure to HCV infection; (3) the outcome was CCA, ICC, or ECC incidence; and (4) providing risk estimates with 95% confidence interval (CI) or available information to calculate them. Abstracts, letters, reviews, case reports, studies lacking control groups, and studies that did not provide sufficient data to calculate risk estimates were excluded. Two investigators independently selected studies, and any discrepancies were resolved by consensus.For each eligible study, the following parameters were extracted independently by two researchers: (1) first author’s last name, date of study publication and the country where the study was conducted; (2) study design; (3) type of control; (4) sample size and age; (5) number of exposure in cases and controls; (6) risk estimate and 95% CI; and (7) adjustments and other information. The present meta-analysis included different measures of risk estimates . In fact, the different measures yielded a similar risk estimate due to the relatively scarce occurrence of cholangiocarcinoma.Q and I2 statistics. For the Q statistic, P < 0.10 was used as an indication of the existence of heterogeneity. Subgroup analysis was performed to explore potential heterogeneity among studies. Publication bias was assessed using the Begg funnel plot and the Egger test, and a P < 0.10 was considered as an indicator for publication bias [A pooled risk estimate was calculated with a random-effects model, considering both intra-study and inter-study variances. Statistical heterogeneity was evaluated using the ion bias ,17. All et al. [et al. [A total of 159 articles were identified through the literature search. Of these, 146 studies were excluded because they were either review articles, case reports, letters, or were not case-control studies. Studies were excluded when exposure was not an HCV infection and outcomes were not CCA, ICC, or ECC. The remaining 13 case-control studies, plus 8 studies from the reference lists, added 21 studies, of which 5 were excluded because of duplicates from the same population and risk estimates were not available. After strict screening, a total of 16 case-control studies were ultimately included in the present study -24 .We identified four case-control studies ,14,20,21P < 0.001) in a random-effects model in the meta-analysis for 13 case-control studies [Q = 64.32, I2 = 81.3%, P < 0.001). In order to explore potential sources of heterogeneity, we conducted subgroup analyses by geographic region, sources, control groups, and adjusted covariates , but not in North American studies. The pooled ORs were 2.24  for studies with hospitalized controls and 3.38  with population-based controls. The heterogeneity was significant only in studies with hospitalized controls .We found a statistically significant positive association between HCV infection and ICC incidence , suggesting that publication bias had little effect on summary estimates.Funnel plots showed no publication bias in studies concerning CCA, ICC, and ECC Figure . P valueThe global prevalence of CCA has shown a rising trend in recent years, and the incidence of HCV infection is also found to be elevated ,25-28. Aet al. [versus 2.01). This phenomenon was also seen in the subgroup analysis of HCV and the risk of ECC (4.88 versus 1.06) in the different geographic regions. This might be explained by the fact that hepatitis C was the most prevalent form of viral hepatitis infection in Western countries compared to Asian countries. As many studies have reported, ICC incidence is rising in Western countries, which may reflect that regional differences have led to different risk factors and epidemiological factors associated with ICC [versus 3.38) and ECC (1.66 versus 1.91).The present study found that patients with an HCV infection have an approximately 5.44-fold increased risk of CCA compared to individuals with a non-HCV infection. In the study by Srivatanakul et al. , the OR with ICC ,31,32. Iet al. [et al. [et al. [et al. [et al. [The mechanism of CCA development in patients with HCV infection has not yet been clarified in detail. One study suggests that the presence of HCV in the bile duct epithelium indicates that HCV infection can cause chronic inflammation of the bile duct epithelium . Long-teet al.  found thet al. . Therefoet al. ,38,39. Set al. ,41, our  [et al.  and Shai [et al. . However [et al.  found th [et al. . Similar [et al.  also fou [et al. . ResultsThis meta-analysis has several limitations that should be considered when interpreting the results. First, conducting randomized clinical trials was not feasible for an observational study of HCV and the risk of CCA. All studies included in this meta-analysis were case-control studies that might introduce selection and recall biases, which may possibly distort the true associations between HCV and CCA. Second, seropositivity for anti-HCV was used as the sole marker of HCV infection, and the definition of HCV infection was not stated in most studies, therefore a common definition may be lacking. Occult HCV infection might also play a role in the development of CCA. Thus, for these reasons, the HCV effect may be underestimated. Third, the diagnostic misclassification of cholangiocarcinoma  might result in a mismatch of true ICC or ECC.The results of the present study reveal that HCV is associated with the significantly increasing risk of CCA, including ICC and ECC, especially ICC. Further studies are warranted to gain a better understanding of the precise mechanism between HCV and CCA.


Objectives: Nonsyndromic cleft lip and palate (NSCLP) is genetically distinct from those with syndromic clefts, and accounts for ~70% of cases with Oral clefts. Folate, or vitamin B9, is an essential nutrient in our diet. Allelic variants in genes involved in the folate pathway might be expected to have an impact on risk of oral clefts. Given the key role of methylenetetrahydrofolate dehydrogenase 1 (MTHFD1) in folate metabolism, it would be of significant interest to assess its role in NSCLP etiology.Study Design: The present study aims at examining the association between MTHFD1 1958G>A polymorphism and NSCLP risk by conducting a case-control study in south Indian population. Our sample comprised of 142 cases with nonsyndromic clefts and 141 controls without clefts or family history of clefting. The MTHFD1 1958G>A polymorphism was genotyped using PCR-RFLP. Results: An increased risk was found for the heterozygous 1958GA  and homozygous 1958AA  genotypes in the children. When the dominant model (AG+AA vs GG) was applied the risk remained the same as co-dominant model, but the level of significance increased . Conclusions: The results indicated the MTHFD1 1958G>A polymorphism to be one of the important genetic determinants of NSCLP risk in South Indian subjects. Key words:MTHFD1, orofacial cleft, SNP, genetics. Nonsyndromic cleft lip and palate (NSCLP) is genetically distinct from those with syndromic clefts, and accounts for ~70% of cases with Oral clefts. The etiology of NSCLP is multifactorial, with both genetic and environmental factors, involving complex gene-gene and gene-environment interactions, and it is these interactions that play a critical role . Folate,implantation, placentation and embryo or organogenesis stages in the manifestation of birth defects. Evidence from epidemiologic studies have conclusively shown that the prenatal folic acid supplementation reduces risk of many congenital anomalies .MTHFD1 is one of the important genes that is involved in folate metabolism. MTHFD1 gene encodes trifunctional enzyme 5,10-methylenetetrahydrofolate dehydrogenase; 5,10-methenyltetrahydrofolate cyclohydrolase and 10-formylotetrahydrofolate synthetase . This enThe MTHFD1, gene is located on chromosome 14q23.3 and spans 71 kb length with a total of 28 exons. Previous studies have reported the association of MTHFD1 gene variants with serum folic acid and homocysteine levels . The com- SubjectsThe sample consisted of 283 individuals ascertained from Cleft and Craniofacial Centre, Sri Ramachandra University, Chennai, India. All cases were evaluated by two different plastic surgeons for their individual phenotypic features, and was also cross verified through their medical records. There is no involvement of oral pathologist in this study. The case group comprised of 142 individuals with NSCLP . The control group was recruited from the same region, included 141 unrelated individuals without clefts or family history of clefting in three generations. The subjects with congenital malformations or major developmental disorders were excluded from the study. The study was approved by the Institutional Ethics Committee of the Sri Ramachandra University, Chennai, India, and all the study subjects gave informed consent. As many of the children were minors, the consent was obtained from their parents or legal guardian.- GenotypingFrom each study subject 3 ml blood sample was collected into an EDTA vacutainer. Genomic DNA was isolated from leukocytes using phenol-chloroform extraction and ethanol precipitation . MTHFD1 - Statistical AnalysisAllele frequencies were calculated by direct counting of alleles. Hardy-Weinberg equilibrium (HWE) analysis was performed by comparison of observed and expected genotype frequencies using chi-squared goodness-of-fit test. The association between different cleft phenotypes and MTHFD1 1958G>A polymorphism was analyzed by χ2-test. To assess the possible association with NSCLP, the comparative analysis for the MTHFD1 1958G>A polymorphism distribution was performed for the whole group of NSCLP (n=142), as well as for the subgroups of patients with CLP (n=123), and CPO (n=19). The same control group (n=141) was considered for all the three case groups. The risk estimates were calculated with the wild type as the reference category. The effect of the 1958G>A variant was examined in 3 hypothetical risk models, co-dominant, dominant (GA+AA vs.GG) and recessive (GA+GG vs.AA) models. The statistical analysis was performed using SPSS statistical software version 16.0  for Windows. A two-sided p-value <0.05 was considered to be statistically significant. From the HapMap populations 20kb up and downstream SNPs around rs2236225 were extracted and LD maps were constructed using Haploview .P=0.726). Significant difference in allele frequencies were found between control and the NSCLP groups ( p=0.004) and homozygous  genotypes. Increased NSCLP risk was also found under dominant  and allelic models . The recessive effect of the 1958A variant (AA vs GG+AG) in cases was not significant . In subgroup analysis, the 1958A variant showed a similar trend of association for CLP group  in our controls was 46.1%, which is analogous to the frequency of 47% found in the GIH population of HapMap (http://hapmap.ncbi.nlm.nih.gov). The frequency reported in the current study is slightly lesser than that (55%) found in a small sample of south Indians . This diAs MTHFD1 is involved in folate pools and folate-dependent reactions, its gene variants have been studied extensively for their association with birth defects in humans ,19,20. TThere is no consistent evidence to show that the MTHFD1 1958G>A polymorphism affected MTHFD1 function or homocysteine levels. The 1958G>A SNP of MTHFD1 results in the substitution of a glutamine (Gln) for arginine (Arg) within the 10-formylTHF synthetase domain of the MTHFD1 enzyme . Under nIn summary, the current study demonstrates a strong association between MTHFD1 1958G>A and risk of NSCLP in south Indian population. Further the relationship between additional mutations in the MTHFD1 gene and serum folate levels or plasma homocysteine levels and their interactions could be examined to provide an eligible conclusion.


The relationship between homocysteine (Hcy) and diabetic retinopathy (DR) remains unclear to date. Therefore, a systematic review and meta-analysis was performed on the relationship between Hcy level and DR.Studies were identified by searching PubMed, Embase, and Web of Science databases until 5 May, 2014.A total of 31 studies involving 6,394 participants were included in the meta-analysis. After pooling the data from each included study, the blood Hcy concentration in the DR group was observed to be higher than that in the control group , and diabetes mellitus (DM) patients with hyperhomocysteinemia were at a risk for DR . Considering the different DM types, hyperhomocysteinemia in T1DM  was associated with DR rather than in T2DM . Considerable statistical heterogeneity in the overall summary estimates was partly explained by the geographical differences.Results from this current meta-analysis indicate that hyperhomocysteinemia is a risk factor for DR, especially proliferative DR. Differences between geographical regions were observed in the relationship between hyperhomocysteinemia with T1DM risk. Given the heterogeneous results, the relationship between high Hcy and DR needs further investigation.http://www.diagnosticpathology.diagnomx.eu/vs/13000_2014_167The virtual slide(s) for this article can be found here:  Diabetic retinopathy (DR), a common complication of diabetes, is the leading cause of low vision and blindness worldwide . To dateHomocysteine (Hcy) is a sulfur-containing amino acid formed by the demethylation of methionine; it is an emerging risk factor for diabetic nephropathy and cardiovascular disease that has gradually elicited the interest of researchers . These rHowever, studies on the relationship between Hcy and DR have reported inconsistent results, and the role of Hcy in the development of DR is not clearly elucidated. Several studies have shown relationships between the blood Hcy level and the prevalence of DR ,6. PlasmGiven the absence of uniform research standards and inconsistency in research results, a meta-analysis of prospective studies on the relationship between Hcy and DR was conducted.This meta-analysis mainly focused on the following objectives: (1) to review and summarize the epidemiologic evidence on the relationship between hyperhomocysteinemia and risk of DR and (2) to quantify the mechanism of hyperhomocysteinemia and DM relationship according to study characteristics.The meta-analysis was conducted according to the Preferred Reporting Items for Systematic Reviews and Meta-Analysis guidelines .We searched PubMed, Embase, and Web of Science to retrieve related studies published before May, 2014. Literature search was performed using keywords “homocysteine,” “hyperhomocysteinemia,” “Hcy,” and “homocysteic acid” in combination with “diabetic retinopathy” and “diabetic retinitis.” Moreover, the citations of related articles were detected for additional publications. No language or date limitations were set. Corresponding authors of the retrieved articles were contacted for additional information.Studies that met the following inclusion criteria were included in this meta-analysis: (1) comparative studies; (2) reported either the prevalence in hyperhomocysteinemia patients and the controls or the concentrations of Hcy in DR and the controls; and (3) provided the value of mean with 95% confidence intervals (CIs), standard errors or sufficient data to calculate them.Studies that met the following inclusion criteria were excluded in this meta-analysis: (1) not for DR research; (2) studies with small sample size (less than 40 cases or control group) and low quality  (3) duplicate of previous publication; and articles without available data; (4) the plasma Hcy levels of studies are influenced seriously by pathological factors.Data were independently extracted from each included article by two reviewers (C. Xu and Y. Wu), and all disagreements on eligibility during the extraction were discussed and resolved by the reviewers. The first author, country, year of publication, number of both case and control groups, blood components used for the test, DM type, definition of hyperhomocysteinemia, and matching factors were extracted.The Newcastle–Ottawa scale (NOS)  was adop2 and I2 tests were used to evaluate the statistical heterogeneity among studies, and the heterogeneity was considered statistical when P < 0.1 and I2 ≥ 50%. Meta-regression method was used to evaluate the source of heterogeneity.This meta-analysis was conducted using the Stata software package . A random-effect method was conducted to pool the results together . The pooThe sensitivity analysis evaluated the robustness of the results in this study. Two different methods were performed to conduct this analysis. First, the articles were excluded one by one, and then a meta-analysis was again conducted. Second, studies with small sample size (less than 40 participants in either case or control group) and low quality  were excluded. To evaluate the potential publication bias, both the funnel plot and the Egger test were performed ,16.Of the 299 articles  identified from database searches, 132 duplicates were excluded. After reviewing the titles and abstract of the 167 remaining articles, 114 articles were excluded and 53 full-text articles were assessed for eligibility. A total of 24 articles (four from duplicated data sources and 20 without available data) were excluded. Moreover, two studies were identified from reference lists. Thirty-one articles were included in this meta-analysis -37, in wUp to 6394 individuals were included in this meta-analysis. A total of 609 cases and 238 controls comprised the comparison between the risk of DR in hyperhomocysteinemia and control groups, and 2070 cases and 3477 controls comprised the comparison between the concentrations of Hcy in DR and NDR groups. The definitions of hyperhomocysteinemia were different in each study, and four of nine studies used 15 μmol/l in the plasma as cut-off. Although the definitions were not the same, no significant differences existed in the definitions. Both plasma and serum were used for the measurement of the concentrations of Hcy in most studies. All of the included articles were retrospective case–control studies. Of the 31 included studies, 4 were conducted in the United States, 11 in Asia, 14 in Europe, and 2 in Oceania. Age and gender were matched in most of the studies. Several studies displayed other comparable factors. Tables NOS was used to evaluate the methodological quality of each study because all of the included articles were retrospective case–control studies. Among the 31 studies, six were less than 6 points, and the mean NOS score was 6.45 ± 1.21 points. The lowest and the highest scales were 4 points in two studies and 8 points in six studies, respectively.2 = 26.9%, P = 0.205). A higher concentration of Hcy in the blood was observed in the DR group . However, a significant heterogeneity was also detected in the comparison .Figure Subgroup analysis was conducted by DM type, geographic site, blood components, and study designs Table . In the In the United States and Asia, a higher incidence rate of DR was detected in the hyperhomocysteinemia group . However, in Europe, the incidence rate of DR was not higher in the hyperhomocysteinemia group compared with the control group . In Asia and Europe, the concentrations of Hcy were higher in the DR group compared with the control group . However, the concentrations of Hcy in the DR groups were not significantly higher in the US and Oceania .Four different study designs, namely, DR vs. normal, DR vs. DM, non-proliferative diabetic retinopathy (NPDR) vs. control, and PDR vs. control, were used in the subgroup analysis. A higher incidence rate in hyperhomocysteinemia was detected in all of the subgroups, and a higher concentration of Hcy was detected in the DR group. No inverse results were detected in the subgroup analysis compared with the main results.2 = 90.9, P < 0.01). However, most subgroup analyses exhibited significant heterogeneity compared with the Hcy concentrations in DR and NDR. No significant heterogeneity was observed between the DR and the NDR groups, whereas the comparison of Hcy serum was significantly heterogeneous .The incidence of DR was compared in the two groups (high Hcy and control). Only the subgroup analysis between the DR and the normal group was significantly heterogeneous  and low quality . After excluding five studies, a significant relationship was observed between hyperhomocysteinemia and risk of DR . Moreover, after excluding 12 studies, the concentration of Hcy in the DR group was still higher than that in the control group .Both the funnel plot Figure  and the A large number of studies focused on the relationship between Hcy and retinopathy, but their results were inconsistent. Studies on T1DM reported positive relationships between DR and Hcy ,39. Hcy In some studies, no significant correlations existed between plasma Hcy and different grades of DR in patients with T1DM, except when a concomitant nephropathy was present ,11,30,36A comprehensive and quantitative summary was lacking on the relationship between Hcy and DR. Therefore, a meta-analysis was conducted to assess the relationship between Hcy and DR. Based on the meta-analysis of the DR groups compared with the normal or DM patients, Hcy concentration in the blood was higher than in the control. These results are the same as those of the PDR groups compared with the control, especially in Asia and Europe. No significant differences were observed in Australia and USA. These findings may be associated with research heterogeneity or the district biases. The different research methods and the sensitivity to detect retinopathy by fundus photography or misclassification of retinopathy may result in district bias.Plasma Hcy levels are influenced by environmental and genetic factors , as wellMany studies excluded serious pathology state of research groups, and different research groups matched for these factors, which could influence plasma Hcy levels. Therefore, we believe that hyperhomocysteinemia may be a potential risk factor for DR patients .2 = 90.9, P < 0.01). This result may due to the different study designs with different methodological qualities.A higher incidence of DR exists in diabetic patients with hyperhomocysteinemia. According to the different DM types, a significant difference in T1DM group is observed, whereas no significant difference is detected in T2DM group. Given all the differences, the incidence of DR in hyperhomocysteinemia cases is higher than in the control, except in Europe. Although most relationships are not significant, a high homocysteinemia is common in the relationship with the incidence of DR, especially with PDR Table . To undeThe pathogenesis of hyperhomocysteinemia in the development of DR may accelerate cell injury (including oxidative stress and impaired generation of nitric oxide) and proliferate smooth muscle cells to alter vasomotor activity ,44. CellA few studies demonstrated that vitreous hyperhomocysteine is related to DR because of the loss of blood–retinal barrier permeability. This result may augment the diffusion of Hcy into the vitreous and the changes in vascular permeability ,48. TherGiven the numerous factors influencing the development and progression of DR, some of them remain unknown. We believe that hyperhomocysteinemia is another contributing factor to DR, especially to the severity of retinopathy independent of other risk factors .The most significant conclusion that can be obtained from the data in Figures Some limitations of this meta-analysis are as follows. First, the included studies are not randomized controlled trials, and thus, may hold potential bias. Second, several outcomes were quite heterogeneous in this meta-analysis, and the subgroup analysis was not successful in detecting the source of heterogeneity. Moreover, we cannot obtain a possible threshold between the blood Hcy level and the separate degrees of DR because of the limited number of subjects with retinopathy and the different study designs.The risk factors for DR are complicated by many aspects. Using Insulin may increase the risk of diabetic retinopathy , HoweverAll these meta-analysis are very significant to show the relationship between the different risk factors with diabetic retinopathy and other diabetic Complications. But our meta show the relationship between Hcy and different aspect of DR deeply.Hcy may be a potential useful biomarker in assessing the microvascular risk in diabetes, and hyperhomocysteinemia is a potential risk factor for DR, especially PDR. This result may help DM patients in obtaining an instructive management to prevent the progression of DR because the plasma Hcy level can be substantially lowered with folic acid supplementation. High Hcy on DR will result in increasing attention to the study of gene C667T polymorphism and related prevention and treatment methods. In addition, considering the limitation of this study, larger studies should be considered and conducted to validate the current meta-analysis.


We investigated the role of the glutathione S-transferase A1, M1, P1 and T1 gene polymorphisms and potential effect modification by occupational exposure to different chemicals in Serbian bladder cancer male patients.A hospital-based case-control study of bladder cancer in men comprised 143 histologically confirmed cases and 114 age-matched male controls. Deletion polymorphism of glutathione S-transferase M1 and T1 was identified by polymerase chain reaction method. Single nucleotide polymorphism of glutathione S-transferase A1 and P1 was identified by restriction fragment length polymorphism method. As a measure of effect size, odds ratio (OR) with corresponding 95% confidence interval (95%CI) was calculated.The glutathione S-transferase A1, T1 and P1 genotypes did not contribute independently toward the risk of bladder cancer, while the glutathione S-transferase M1-null genotype was overrepresented among cases . The most pronounced effect regarding occupational exposure to solvents and glutathione S-transferase genotype on bladder cancer risk was observed for the low activity glutathione S-transferase A1 genotype . The glutathione S-transferase M1-null genotype also enhanced the risk of bladder cancer among subjects exposed to solvents . The risk of bladder cancer development was 5.3–fold elevated among glutathione S-transferase T1-active patients exposed to solvents in comparison with glutathione S-transferase T1-active unexposed patients . Moreover, men with glutathione S-transferase T1-active genotype exposed to pesticides exhibited 4.5 times higher risk in comparison with unexposed glutathione S-transferase T1-active subjects .Null or low-activity genotypes of the glutathione S-transferase A1, T1, and P1 did not contribute independently towards the risk of bladder cancer in males. However, in association with occupational exposure, low activity glutathione S-transferase A1 and glutathione S-transferase M1-null as well as glutathione S-transferase T1-active genotypes increase individual susceptibility to bladder cancer. Bladder cancer is the second most common malignancy of the urinary tract and has the second highest mortality rate among urological neoplasms The well established risk factors for bladder cancer include cigarette smoking , but also exposure to occupational agents GSTM1, GSTT1 and GSTP1 genes receiving the most attention in genetic epidemiological studies GSTM1*0 and GSTT1*0 alleles (frequently referred to as GSTM1-null and GSTT1-null genotypes), which comprise for 50% and 11–18% of white population, respectively GSTP Ile/Ile, -Ile/Val and -Val/Val are 51.5, 39.4, and 9.1%, respectively GSTA1 polymorphism has emerged relatively recently in genetic epidemiological studies. It is represented by three, apparently linked, single nucleotide polymorphisms (SNPs): -567TOG, -69COT, -52GOA GSTA1*B  than common GSTA1*A allele  GSTA1-AA, AB and BB genotype in Caucasians are 38%, 48% and 14%, respectively Despite the fact that occupations associated with bladder cancer have been well established, the question still arises why individuals with seemingly equal exposure to occupational carcinogens develop bladder cancer in an unpredictable manner. This is probably attributed to genetic polymorphisms of the genes coding for the xenobiotic metabolizing enzymes, particularly glutathione S-transferase (GST). GSTs catalyze the conjugation of glutathione on electrophilic substrates and are an important line of defense in the protection of cellular components against reactive species. The most well characterized GST classes have been named alpha (GSTA), mu (GSTM), pi (GSTP) and theta (GSTT). Appreciable GST activities are seen in bladder epithelium GST gene variants coding for enzymes involved in biotransformation of specific occupational agents may influence the risk of occupational bladder cancer. Therefore, in this case-control study we investigated the role of the polymorphisms GSTA1, GSTM1, GSTP1 and GSTT1 gene and potential of effect modification by occupational exposure to different chemicals in Serbian male TCC patients.Many of the well known occupational agents, such as polycyclic aromatic hydrocarbons, aromatic amines, halogenated hydrocarbons, associated with bladder cancer risk are substrates for GST. Although this reaction generally results in detoxification, in selected cases GST-mediated conjugation may lead to a more toxic or mutagenic metabolite. Still the data on association between GST gene variants and risk of occupational bladder cancer are scarce. We hypothesized that This study was approved by the Ethical Committee of Faculty of Medicine, University of Belgrade and conducted according to the principles expressed in the Declaration of Helsinki. All the participants provided written informed consent.A hospital-based case-control study of urinary bladder cancer in men was carried out between September 2007 and January 2010. A total of 143 histologically confirmed incident urinary bladder carcinoma male cases were recruited from the Clinics of Urology and Nephrology, Clinical centre of Serbia, Belgrade. This is the national reference center for urology and nephrology and the majority of bladder cancer patients from Serbia are diagnosed and treated at this clinic. The control group consisted of 114 male subjects which were recruited from individuals with nephrolithiasis admitted to the same hospital during the same period of time and had no history of any malignant disease. Urinary bladder carcinoma patients and corresponding controls did not differ with respect to mean age .After the informed consent was obtained, each subject was interviewed by well-trained interviewers using a standard questionnaire to collect information including demographic characteristics, history of cigarette smoking and occupational exposure. Response rate was 92% and the most frequent reason for no participation was personal.In our study, smokers were defined as persons who reported every day smoking during a minimum of 60-day period prior to completing the questionnaire. Participants were asked about the number of cigarettes smoked per day and duration of smoking. The amount of pack-years was calculated using the following formula: pack-years  =  (cigarettes/day÷20)×(smoked years).The life-time occupational history listed all jobs  lasting more than six months and consisted of the job title, the industry or type of business, employment dates and duration, company name and location, tasks as well as the exposure to at least one of the categories of agents under study, solvents and pesticides. In order to analyze occupational exposure occupational reports of patients were evaluated by experienced specialist in occupational medicine . The exposure categories were defined as no exposure and exposure. Based on the evaluation patients were exposed to the following organic solvents: tetrachloroethylene, toluene, xylene, ethyl acetate, acetone, petrol ether and ethanol, as well as pesticides: organophosphate, carbamates, aminophosphonic analogues, chloroacetanilides, derivative of benzoic acid. All exposure data referred to a time period prior to the diagnosis of bladder cancer for the cases, and a corresponding period for the controls.Qiagen, Inc., Chatsworth, CA, USA) 96-spin blood protocol according to the manufacturer's instructions. Blood was collected when patients were admitted to the clinic.Genomic DNA was isolated from whole blood using the QIAGEN QIAmp  PCR product corresponded to 312 bp.GSTT1 genotyping was performed by multiplex PCR method GSTT1-forward: 5′-TTCCTTACTGGTCCTCACATCTC-3′ and GSTT1-reverse: 5′-TCACGGGATCATGGCCAGCA-3′. Exon 7 of CYP1A1 genes were co-amplified and used as an internal control. The assay does not distinguish between heterozygous or homozygous wild-type genotypes; therefore, the presence of 480 bp bands was indicative for the GSTT1-active genotype. Internal positive control (CYP1A1) PCR product corresponded to 312 bp.GSTP1 Ile105Val polymorphism was analyzed using the polymerase chain reaction–restriction fragment length polymorphism (PCR–RFLP) GSTP1 Ile105Val forward: 5′-ACCCCAGGGCTCTATGGGAA-3′ and GSTP1 Ile105Val reverse: 5′-TGAGGGCACAAGAAGCCCCT-3′. The amplification 176 bp products (20 µl) were digested by 10 U of restriction endonuclease Alw261at 37°C over night. The presence of restriction site resulting in two fragments (91 and 85 bp) indicated mutant allele , while if Ile/Val polymorphism incurred, it resulted in one more fragment of 176 bp.GSTA1 C-69T polymorphism was determined by polymerase chain reaction–restriction fragment length polymorphism (PCR–RFLP) according to Coles et al GSTA1 C-69T forward:5′-TGTTGATTGTTTGCCTGAAATT-3′ and GSTA1 C-69T reverse: 5′-GTTAAACGCTGTCACCCGTCCT-3′. The amplification 481 bp products (20 µl) were digested by 10 U of restriction endonuclease Ear1 at 37°C over night. The presence of restriction site resulting in two fragments (385 and 96 bp) indicated mutant allele (B/B) and if A/B polymorphism incurred, it resulted in one more fragment of 481 bp.All genotyping was performed by laboratory personnel blinded to case-control status, and blinded quality control samples were inserted to validate genotyping identification procedures; concordance for blinded samples was 100%.GSTA1 and GSTP1 polymorphisms for the case and control populations was tested for the Hardy–Weinberg equilibrium by χ2 test. As a measure of effect size, odds ratio (OR) with corresponding 95% confidence interval (95%CI) was used to describe the strength of association between the genotypes and bladder cancer modified by occupational exposure. Unconditional logistic regression analysis is applied. Bearing in mind that age and smoking are well established risk factors for bladder cancer, we adjusted OR by these variables as potential confounders. Interactions between GST polymorphisms and occupational exposure were included in the logistic regression models and also adjusted by potential confounding variables. The probability level of ≤0.05 was considered statistically significant. For statistical analysis the SPSS 17.0 statistical software package  was used.The distribution of the GSTA1 and GSTP1 genotype frequencies were in Hardy-Weinberg equilibrium both for cases and controls (p>0.05). The observed genotype frequencies in controls were not significantly different from frequencies previously described among Caucasians. However, the frequency of GSTT1-null genotype in control group (28%) was higher than values reported among Caucasians (18.1%). As shown in GST null/low-activity genotypes were higher in cases compared to controls with the exception of the GSTT1-null genotype. Although GST A1, T1 and P1 genotypes did not contribute independently toward the risk of TCC, the GSTM1-null genotype was overrepresented among cases (56%) compared to M1-active genotype with an adjusted OR of 2.1 reaching a statistical significance .Genotyping was conducted for all recruited patients . The GSTGSTA1, GSTM1, GSTP1 and GSTT1 polymorphisms and occupational exposure on bladder cancer risk in male patients are shown in GST genotype. We found that occupationally exposed individuals with GSTT1-active genotype exhibited 4.3-fold increased risk compared to the unexposed T1-active subjects . However, only for the GSTP1 gene is there evidence of a gene–occupational exposure interaction (p = 0.017).Combined effects of GST genotype on bladder cancer risk in male patients is shown on GST polymorphisms tested. The most pronounced effect regarding occupational exposure to solvents and GST genotype on bladder cancer risk was observed for the GSTA1 genotype, since men exposed to solvents with GSTA1-low activity genotype had 9 times higher risk of bladder cancer than GSTA1-active unexposed men . Similarly to that observed for GSTA1-low activity, the GSTM1-null genotype enhanced the risk of TCC among subjects exposed to solvents compared to the unexposed GSTM1-active individuals . These results point to the importance of antioxidant GSTA1 and GSTM1 activity protection against free radicals produced during solvent metabolism. The risk of TCC development was 5.3–fold elevated among GSTT1-active patients exposed to solvents in comparison with GSTT1-active unexposed patients . Significant association was also found for GSTP1 Ile/Ile individuals who had 3.3 higher TCC risk compared to the unexposed Ile/Ile individuals . However, only for GSTP1 statistically significant interaction between genotype and occupational exposure to solvents was found (p = 0.044)In order to test whether GST-occupational exposure interaction is modified by the specific type of exposure, cases and controls were further stratified into exposed to solvents and exposed to pesticides. Combined effect of occupational exposure to solvents and GST genotype on bladder cancer risk in male patients is shown on GSTT1-active genotype exposed to pesticides exhibited 4.5 times higher risk in comparison with unexposed GSTT1-active subjects .Combined effect of occupational exposure to pesticides and Our results showed that occupationally exposed men had 3 times higher risk for TCC. This result confirms the occupational exposure as a TCC risk factor low activity GSTA1 allele had the highest risk (about 9 times), while GSTM1-null carriers had 6.5 times higher bladder cancer risk when compared to unexposed GSTA1 AA and GSTM1-active persons, respectively. This result was expected since in several malignant diseases, such as colorectal, prostate and hepatocellular cancer, GSTA1*B allele with lower transcriptional activity was associated with increased risk. GSTA1 protein belongs to the most promiscuous GSTs that acts upon a broad range of substrates which bind to its active site low-activity GSTA1 and GSTM1-null genotype increase susceptibility to bladder cancer in occupationally exposed men can be explained by the role of GST enzymes in detoxification and in antioxidant defense. Namely, GSTA1 and GSTM1 possess strong peroxidase activity and are key components in cellular defense against free radicals GSTT1-active individuals occupationally exposed to solvents exhibited 5 times higher risk of TCC in comparison with GSTT1-active unexposed subjects. These results are biologically plausible since GST-mediated conjugation with halogenated substrates may lead to a more toxic or mutagenic metabolites. Namely, substrates with ≥2 halogenes are activated because the conjugated product is instable, leading to reactions with nucleophiles, particularly DNA and proteins GST genes produce enzymes that metabolize some solvents to cytotoxic metabolites; while those with less functional or nonfunctioning genes have little or no enzyme and apparently do not produce cytotoxic metabolites from solvent exposure. Until now, the association between GST polymorphism and occupationally related cancers has been studied mostly in renal cell carcinoma. Results of these studies showed that GSTT1-active genotype enhanced the risk of renal cell carcinoma among subjects exposed to solvents. Our results on higher bladder carcinoma risk in GSTT1-active individuals occupationaly exposed to solvents are in accordance with previously published results in renal cell carcinoma It is important to note that risk imposed by occupational hazards was modified by GST polymorphism. We observed that individuals occupationally exposed to solvents with at least one GSTM1-active genotype GSTT1-active individuals exposed to pesticides when compared to GSTT1-null workers similarly exposed GST polymorphism and occupational exposure to pesticide with respect to risk of carcinoma of urinary tract. Namely, Karami and others reported that renal cell carcinoma risk associated with pesticide exposure was highest among individuals with active GSTM1/T1 genotypes GSTT1-active genotype. One of the reasons for non-significant association between GSTT1-active genotype may be the relatively small number of pesticide exposed participants in both case and control groups. Nevertheless, it is well known that pesticides produced from halogenated alkanes, alkenes undergo bioactivation in the liver and kidney after conjugation to glutathione by GSTT1 GSTT1-genotype will form an inactive enzyme and therefore metabolism of halogenated compounds will occur through oxidation, without formation of reactive intermediates Although it has been postulated that exposure to pesticides and/or fertilizers might be responsible for higher urinary bladder risk, the evidence is still conflicting. Some studies have shown that TCC risk was significantly elevated among men in the landscape and horticultural services industry, as well as in gardeners, and lawn care service employees GST polymorphisms studied might be sources of potential biases which may influence the study findings. However, we tested effects of four GST polymorphisms and occupational exposure on bladder cancer risk and therefore significantly decreased chance for publication bias. Additionally, we cannot entirely rule out the possibility that some of our results could be caused by confounding, although we included only men and adjusted all results by age and smoking status. Further studies with larger samples and more rigorous designs are needed to investigate the gene effects and the potential effect modification by environmental factors.The principal limitations of this study are the relatively small sample size which limiting the precision of the odds ratios, hospital-based control group and qualitative evaluation of occupational exposure. Concerning the actual sample size (143 cases and 114 controls), the statistical power is 66%. Furthermore, it is well known that relatively small numbers of both study participants and GSTM1-null genotype increased the risk of bladder cancer in males. Null or low-activity genotypes of the GSTA1, GSTT1, and GSTP1 did not contribute independently towards the risk of bladder cancer in males. However, in association with occupational exposure, both low activity GSTA1 and GSTM1-null genotype increase individual susceptibility to bladder cancer suggesting the protective role of these detoxification and antioxidant enzymes in metabolism of occupational hazards, specifically organic solvents. On the other hand, the presence of GSTT1-active genotype in occupationally exposed subjects, resulting in GSTT1 protein expression and GSTT1 mediated bioactivation, increases the risk of bladder cancer.