The genetic polymorphisms of glutathione S-transferase (GSTs) have been suspected to be related to the development of lung cancer while the current results are conflicting, especially in the Chinese population.GSTM1) from 68 studies, glutathione S-transferase theta 1 (GSTT1) from 17 studies and GSTM1-GSTT1 from 8 studies in the Chinese population were reanalyzed on their association with lung cancer risk. Odds ratios (OR) were pooled using forest plots. 9 subgroups were all or partly performed in the subgroup analyses. The Galbraith plot was used to identify the heterogeneous records. Potential publication biases were detected by Begg's and Egger's tests.Data on genetic polymorphisms of glutathione S-transferase Mu 1 .71 eligible studies were identified after screening of 1608 articles. The increased association between two vital GSTs genetic polymorphisms and lung cancer risk was detected by random-effects model based on a comparable heterogeneity. Subgroup analysis showed a significant relationship between squamous carcinoma (SC), adenocarcinoma (AC) or small cell lung carcinoma (SCLC) and GSTM1 and GSTT1 null genotype, especially with dual null genotype, was found in the Chinese population. In addition, special histopathological classification of lung cancers and a wide range of gene-environment and gene-gene interaction analysis should be taken into consideration in future studies.An increased risk of lung cancer with  Lung cancer is the most common malignancy in the world and the leading cancer in males, accounting for 17% of the total new cancer cases and 23% of the total cancer deaths Lung cancer can be divided into several types by pathological classification, such as squamous cell carcinoma (SC), adenocarcinoma (AC) and large or small cell carcinoma. It is also classified as small cell lung carcinoma (SCLC) and non-small cell lung carcinoma (NSCLC), which accounts for about 85% of all lung cancer 2−) and hydroxyl radicals (OH−) and was repaired by antioxidant enzymes. This balance can be broken by both environmental and genetic factors. Available molecular epidemiology studies have shown that genetic polymorphisms play a major role in the progress of carcinoma Tobacco smoking has clearly been demonstrated to be a strong exogenous factor for lung cancer risk GSTM1 and GSTT1 are two critical GSTs family genes, separately encoded mu and theta GST classes and located in 1p13.3 and 22q11.23 in the human chromosome, respectively. The common GSTM1 polymorphisms include three alleles, GSTM1*A, GSTM1*B and GSTM1*0, where GSTM1*0 means a null mutation GSTT1 is polymorphic with two alleles (GSTT1*1 and GSTT1*0). The homozygous combinations of GSTM1*0 allele as a null genotype could lead to a functional deficiency GSTT1*0The GSTs family can detoxify environmental carcinogens and toxins, oxidative stress products, and several covalent conjugated electrophilic compounds GSTM1 is observed in the Asian population Most molecular epidemiologic studies suggested an association between GST genetic polymorphisms and lung cancer risk, especially when deletion of GSTM1 and GSTT1) with lung cancer risk, an updated systematic meta-analysis was performed in this study by selecting all eligible studies in the Chinese population.To identify the association of two vital GST genetic polymorphisms  to identify additional eligible studies. MeSH terms (“glutathione S-transferase” or “GST” or “GSTM1” or “GSTT1”) and (“lung carcinoma” or “lung cancer” or “lung neoplasms”) and (“China” or “Chinese” or “Taiwan”) were used in the databases. Eligible research articles not captured by the above research strategies were further searched by bibliographies without language limitation.A computer-based literature search was carried out in EMBASE, PubMed, ISI Web of Knowledge, Chinese Biomedical Database (CBM), VIP database, Chinese National Knowledge Infrastructure (CNKI), and Wanfang Data (the latest research retrospect until October 2013) to collect articles related to the association of Inclusion criteria: (1) individuals or samples in all eligible studies were examined and diagnosed by polymerase chain reaction (PCR), pathologic diagnosis or other methods to get a full picture of GST genetic polymorphisms and lung cancer types; (2) Chinese living in China; (3) articles providing raw data including odds ratio (OR) with 95% confidence interval (CI) and respective variance, or the relevant information could be calculated.Exclusion criteria: (1) Chinese out of China; (2) raw data not available; (3) when there were multiple publications by the same researchers, only the latest or the largest population study was adopted; (4) meeting abstract, case reports, editorials, newsletter and review articles were excluded.GSTM1, null GSTT1, and null of both genotypes  using a standardized data extraction form designed by ourselves. Discrepancies among three groups were further discussed by all parties. If consenses was still not reached, another group (group 4-Huiqin Wang and Qi Zhou) would make the final decision. Firstly, the titles and abstracts of all studied articles were screened to determine their relevance. If the titles and abstracts were ambiguous, full articles would be investigated. In order to make full use of the available data, it was counted as two separated studies if two different control groups were employed in the same article, such as two different controls versus the same control. If there were more than one region to be investigated in one article, information for each region was also counted as a separated study. Information collected from each eligible study included: first author, year of publication, region, study time, pathologic diagnosis, source of control, characteristics of cases and controls, genotype frequency of null enotypes . Hardy Welicited . Study tP≤0.05 was considered statistically significant. (2) Statistical heterogeneity among studies was assessed by Q and 2I statistics P≤0.1, a random-effects model was used; when P>0.1, a fixed-effects model was performed 2I≥50%, 50%>2I≥25% or 2I<25%, we identified the studies as high, middle or low heterogeneity, respectively. (3) Sensitivity analysis was performed by removing one study at a time to calculate the overall homogeneity and effect size; the Galbraith plot was also performed to examine the possible distinct articles. (4) The possible reasons for heterogeneity between studies were investigated by subgroup analyses. Nine subgroups were analyzed as follows: histopathological classification , geographical location  (See vs. non-smoker), CYP1A1(Msp1) polymorphisms, case number (<100 vs. ≥100), source of controls , research design , test material  and quality score . The last five items listed above were used to assess the study quality. (5) Cumulative meta-analysis was used to explore any significant changes in the variation of sample size or publication year. (6) Publication bias was investigated by the Begg's test P values were two sided.(1) The pooled ORs and 95% CIs were determined by the Z test, ina) See , smokingGSTM1 and/or GSTT1 genetic polymorphisms and lung cancer risk from both Chinese and English databases (GSTM1 (8649 cases and 10380 controls) GSTT1 (2109 cases and 3031 controls) GSTM1 and GSTT1 (775 cases and 1495 controls) We ultimately identified a total of 71 articles atabases . There wGSTM1 studies, 13/17 of GSTT1, and 5/8 of both GSTM1 and GSTT1). According to our criterion, 61 (89.7%) studies of GSTM1, 13 (76.5%) of GSTT1, and 7 (87.5%) of both GSTM1 and GSTT1 were evaluated as epidemiological designs. In both control and case groups, 50 (73.5%) studies of GSTM1, 13 (76.5%) of GSTT1 and 7 (87.5%) of both GSTM1 and GSTT1 used white blood cells for GSTs genotype detection. The rest of the studies used adjacent lung tissue, tumor tissue, BALF cells or buccal cells, etc., for GSTs genotype detection in cases or controls. Only two studies reported the HWE test results for the GSTM1 or GSTT1 and satisfied HWE GSTM1 and GSTT1 ranged from 29.7% to 67.9% (Mean = 49.5%) and 37.5% to 63.0% (Median = 44.4%), respectively. The CYP1A1 (Msp1) polymorphisms satisfied the HWE in the controls of 15 (68%) studies about GSTM1 and CYP1A1 (Msp1). More details are shown in Most studies were published in Chinese    (The results showed a significant association between the P<0.001) . The ranP<0.001) . Further= 0.011) . Risk esGSTM1 tended to be stable. However, no significant difference in the trend was found in the GSTT1 and GSTM1-GSTT1 cumulative meta-analysis. The results for cumulative meta-analysis are shown in The cumulative meta-analysis was used to examine the fluctuation of the eligible studies with changes in the publication year or sample size. With the publication year development and sample size increase, the cumulative meta-analysis of GSTM1 in the northwest, or null genotype of GSTT1 in the north, southwest or northwest of China  with mt/mt genotype and GSTM1 null genotype could enhance the risk of lung cancer, and the OR of which were a little higher than the other two CYP1A1 (Msp1) genotypes with GSTM1 null.Due to the fact that all studies were middle to high heterogeneities, analyses on nine subgroups as mentioned above were performed accordingly. No significant increase in the risk of lung cancer was detected in either null genotype of of China , Table 4GSTM1 variants and lung cancer were found in the studies from northeast and southwest China. The subgroups of AC and smokers also showed greater heterogeneities . Meanwhile, the subgroup analyses of GSTT1 genetic polymorphisms and lung cancer susceptibility demonstrated high heterogeneities in the subgroups of central China, southwest China, and smokers.However, high heterogeneities in the analysis of the association between GSTT1 genetic polymorphisms and lung cancer risk, but high heterogeneities also appeared. However, subgroups of case numbers <100, hospital-based controls, non-epidemiological studies, test materials from involved tissue or cells or not available, and quality score (4–5), all pooled results showed no significant association between GSTT1 genetic polymorphisms and lung cancer risk , all pooled results showed significant association between cer risk .GSTM1-GSTT1 genetic polymorphisms with lung cancer risk, no significant association was found in the subgroup of case numbers (≥100). Along with significant increase risks in the subgroup of population-based controls and epidemiological studies, high heterogeneity was also found . Besides, according to the Galbraith plot of the association of GSTT1 or GSTM1-GSTT1 interaction polymorphisms with lung cancer risk, 2 articles GSTT1 = 1.18, 95% CI: 1.10 to 1.26, P<0.001; ORGSTM1-GSTT1 = 1.33, 95% CI: 1.10 to 1.61, P = 0.004) and the 2I indexes were decreased to 29.5% for GSTT1 and 2.1% for GSTM1-GSTT1, respectively . No publication bias was detected by Egger's test . The Trim and Fill test further confirmed the results (data not shown).Begg's funnel plots and Egger's linear regression test were used to evaluate the potential publication bias  and SCLC  risk were found with a low heterogeneity. These results for the first time imply a clue that SCLC could have a stronger association with GSTM1 deficiency than the other two types while no statistic difference was found among 3 pathological types from available data. Due to the limited number of studies and comparatively diversity among various studies, more well designed epidemiological studies should be performed for various pathological types of lung cancers . Additionally, we found that there was increased susceptibility between GSTM1 null genotype and lung carcinoma risk in different phase I isoenzymes of CYP1A1. These results not only further confirm our conclusion, but also imply some enlightenments. For instance, under a higher OR with no heterogeneity, people with CYP1A1 (mt/mt) and GSTM1 null genotype should pay more attention to avoiding exposure to harmful environmental factors associated with lung cancer. Naturally, more studies including a genome-wide association study (GWAS) are necessary to prove this hypothesis. Due to the limited number of studies, the same analysis for the GSTT1 null genotype was not performed.In subgroup analysis of GSTM1 studies further suggested that the possible risk factor of GSTM1 null genotype is different. However, eligible studies for GSTT1 failed to reach a significant association, which might be caused by a limited number of studies with high heterogeneities. Unclear smoking definition and inconsistent classification of the amount of tobacco consumed among different studies might all have an influence on the stability, reliability, as well as further in-depth analyses of the results. Therefore, clear smoking definition and consistent classification for the smoking status are necessary in any future research.The subgroup analyses of the smoking status for GSTM1 were detected by the Galbraith plot. The potential bias of these articles might be the result of small sample size, complex population composition, distinction of testing materials GSTT1 and GSTM1-GSTT1 groups spotted two of the same articles GSTT1 and GSTM1-GSTT1 were both increased.In the sensitivity analyses and Galbraith plot, 7 heterogeneous articles for GSTT1 or GSTM1-GSTT1 with the publication time development and sample size increase. Thus, to identify the real association between the GSTT1 null type, GSTM1-GSTT1 dual null type and lung cancer susceptibility, more large-scale case-control and cohort studies from multi-centers should be performed. At last, no publication biases were detected in our meta-analysis.Cumulative meta-analysis showed a comparable change in the trend in the accumulated OR and 95% CI for GSTM1 and GSTT1 in the present genotype, so Hardy-Weinberg equilibrium tests could not be performed. Therefore, the Hardy-Weinberg equilibrium test results reported in some of the 71 articles might not be reliable.It's worth mentioning that Hardy-Weinberg equilibrium has been widely recommended in testing studies of genetic polymorphisms and diseases, the violations of which may have potential impacts on the results It is worthy to note that several other limitations might be included in this study: (1) as common observational studies, case-control studies were susceptible to various biases . These biases could influence the final findings of this study; (2) conclusions of this study were partly based on literatures obtained from the hospital-based population, which might not represent the whole population; (3) eligible studies for this study covered nearly all regions in China, but the article number was still insufficient in some less developed or relatively sparsely regions; (4) the interaction of genes with environmental factors, especially with special external occupational exposure and environmental pollution, might all contribute to the development of lung cancer. Factors above might also contribute to a possible source of heterogeneity of our results. Owning to the limitation of the data, this paper did not analyze the interaction effects of these factors; (5) absence of HWE test in the control group, some unbalance controls could lead to some bias in the final results.GSTM1 and GSTT1 genetic polymorphisms are associated with increased lung cancer risk in the Chinese population. Because of multifactor etiology of the interaction of gene-gene and gene-environment in the development of lung cancer, large-scale and methodologically sound studies with different environmental background and other genetic polymorphisms should be carried out to explore the real association between GSTs variants and various pathological types of lung cancer.Taken together, after a decade of extensive studying on this topic, our findings suggest that Checklist S1PRISMA checklist.(DOC)Click here for additional data file.Figure S1Map of the seven regions in China.(TIF)Click here for additional data file.


Recent epidemiological evidence points to an association between gallstones or cholecystectomy and the incidence risk of liver cancer, but the results are inconsistent. We present a meta-analysis of observational studies to explore this association.2.IWe identified studies by a literature search of PubMed, EMBASE, Cochrane Central Register of Controlled Trials, and relevant conference proceedings up to March 2014. A random-effects model was used to generate pooled multivariable adjusted odds ratios (ORs) and 95% confidence intervals (CIs). Between-study heterogeneity was assessed using Cochran’s Q statistic and the n = 11 studies), as well as cholecystectomy , but there was considerable heterogeneity among these studies. The effects estimates did not vary markedly when stratified by gender, study design, study region, and study quality. The multivariate meta-regression analysis suggested that study region and study quality appeared to explain the heterogeneity observed in the cholecystectomy analysis.Fifteen studies (five case-control and 10 cohort studies) were included in this analysis. There were 4,487,662 subjects in total, 17,945 diagnoses of liver cancer, 328,420 exposed to gallstones, and 884,507 exposed to cholecystectomy. Pooled results indicated a significant increased risk of liver cancer in patients with a history of gallstones (OR = 2.54; 95% CI, 1.71–3.79; Our results suggest that individuals with a history of gallstones and cholecystectomy may have an increased risk of liver cancer. Primary liver cancer mainly includes hepatocellular carcinoma (HCC), which originates in liver cells, and intrahepatic cholangiocarcinoma (ICC), which arises from the intrahepatic bile duct It has been hypothesized that gallstones  and cholecystectomy are associated with an increased risk of several cancers, especially the risk of rectal cancer Several epidemiological studies have investigated the association between gallstones, cholecystectomy, and liver cancer No meta-analysis has previously been published on the relationship between gallstones or cholecystectomy and the incidence risk of liver cancer. The aim of this detailed meta-analysis was to summarize the association between cholecystectomy, gallstones, and the risk of developing liver cancer in observational studies. A better understanding of these relationships may highlight the need to consider additional intervention methods in this area.This study complies with the guidelines of the PRISMA  checklist and flow diagram American College of Gastroenterology, American Association for the Study of Liver Diseases, Digestive Diseases Week; World Congress of the International Hepato-Pancreato-Biliary Association) between 2009 and 2013. The reference lists in all identified articles were checked for further relevant articles.We searched PubMed, EMBASE, and Cochrane Central Register of Controlled Trials (CENTRAL) in the Cochrane Library for all relevant articles on the risk of liver cancer in patients with history of gallstones or cholecystectomy. The search was performed in each database from time of inception until March 12, 2014 by two independent investigators (Y.L. and S.L.). Medical subject heading terms and keywords used in the search included “cholecystectomy”, “gallbladder surgery”, “gallstones”, “gallstone”, “cholelithiasis”, “cholecystolithiasis”, “choledocholithiasis” combined with “HCC”, “hepatocellular carcinoma”, “liver cancer”, “liver tumors”, “liver neoplasms”, “hepatic carcinoma”. No language restrictions were imposed. We also reviewed the abstracts submitted to major gastroenterology and hepatology conferences  cohort or case-control; (2) focus of the study was a history of gallstones or cholecystectomy; (3) end point was liver cancer incidence; (4) provided multivariate-adjusted relative risks, with corresponding 95% confidence intervals (CIs) for events associated with gallstones or cholecystectomy vs. controls, at least adjusted for three of eight factors . Primary exclusion criteria were cross-sectional studies, literature reviews, commentaries, editorials, and case reports. Studies were also excluded where adjusted relative risks and/or CIs had not been provided, or they did not fulfill the inclusion criteria. When there were multiple studies of the same population, only data from the most recent comprehensive report was included. Two authors (Y.L. and Y.H.) independently evaluated all records by title and abstract and subsequently retrieved and assessed, in detail, the full text of any potentially relevant articles using the above eligibility criteria. Disagreements regarding eligibility were resolved through discussion and by referencing the original report.Data were independently abstracted onto a standardized form by two investigators (T.L. and L.X.). Disagreements were resolved through consensus, referring back to the original article. The following data were collected from each study: first author’s name, year of publication, country of the population studied, mean age, study duration, number of patients with gallstones or cholecystectomy studied, number of incident cases of liver cancer, adjustment factors, and multivariable adjusted relative risk estimates and their 95% CIs . Because the liver cancer incidence is low (≤10%) and the estimated effects are small, odds ratios (ORs) can be considered close approximations of risk ratios The methodological quality of the case-control and cohort studies was initially assessed independently by two authors using the Newcastle-Ottawa scale (NOS) The primary analysis focused on assessing the risk of liver cancer in patients with a history of gallstones, and risk of liver cancer in patients with a history of cholecystectomy. Additionally, based on information available from individual studies, we assessed sex-specific differences in risk estimates.P-value <0.10 was considered suggestive of significant heterogeneity 2I statistic was calculated 2I less than 30% was considered as low, 30%–60% as moderate, 60%–75% as substantial, and more than 75% as considerable We used the random-effects model described by DerSimonian and Laird To explore sources of heterogeneity between the combined studies, we performed subgroup analyses based on study design (case-control vs. cohort), study location , and study quality . In addition, a restricted maximum likelihood-based random-effects meta-regression analysis was performed to assess heterogeneity associated with the aforementioned factors P-value less than 0.05 was considered statistically significant for all analyses except for the Cochran’s Q test. All analyses were performed using STATA, version 12.0 .Sensitivity analyses were conducted to assess the robustness of the results by sequential omission of individual studies n = 8) were conducted in European populations 2I = 97.8%; P<0.001).Eleven studies Some studies We then conducted further subgroup meta-analysis according to study design, study location, and study quality . No subsP = 0.200 for study design; P = 0.892 for study location; P = 0.905 for study quality). The meta-regression analysis indicated that study design, study location, and study quality might not be major sources contributing to the heterogeneity presented in the overall analyses.We also conducted a meta-regression analysis to investigate the impact of heterogeneous factors on the OR estimates. The study design, study location, and study quality were chosen as the potential heterogeneous factors. In multivariate meta-regression analysis, none of these factors were significant  when the study by Welzel et al. P value for the Egger test was 0.847, suggesting a very low probability of publication bias.In the publication bias assessment, the inverted funnel plot appeared to be symmetric . The P vP<0.001, 2I = 91.0%) (Twelve eligible studies = 91.0%) .Four studies reported sex-specific risk estimates of liver cancer When stratifying the data into subgroups based on study design, study location and study quality , we founP = 0.039 and P = 0.008, respectively), with these two variables explaining most between-study variability.We then conducted a meta-regression analysis to investigate the impact of study design, study location, and study quality on the estimated ORs. In multivariate meta-regression analysis, study region and study quality were significant factors  when the study by Welzel et al. P-value of Egger’s regression test (P = 0.581) was more than 0.05, indicating no statistical evidence of publication bias . In the current study, cholecystectomy was associated with a 62% excess risk of liver cancer. The association persisted across a broad range of sensitivity analyses. Further, the significant association was observed in both women and men. However, there was considerable heterogeneity among most analyses.There is a long-standing debate about the risk of cancer in patients who have gallstones and undergo cholecystectomy. Several review studies have discussed the potential risk of gallstones or cholecystectomy in various tumors, such as colorectal The pathophysiology of tumorigenesis associated with gallstones and after cholecystectomy has yet to be elucidated. One potential mechanism may involve chronic inflammation. Gallstones may induce biliary inflammation, and cholecystectomy is typically followed by dilation of the common bile ducts and elevated bile duct pressure P = 0.794, 2I = 0.0%) and low-quality studies . To further investigate the heterogeneity, multivariate meta-regression analyses were performed. Meta-regression analysis of the data showed that the study region and study quality might substantially influence the initial heterogeneity. When these two parameters were considered together, the results indicated that the U.S. and Asian studies might be a major source of the heterogeneity in the cholecystectomy data.Heterogeneity is a potential problem when interpreting the results of all meta-analyses, and finding the sources of heterogeneity is one of the most important goals of any meta-analysis The presence of heterogeneity in study design, population characteristics, sample size, information collection methods , definition of exposure, assessment of outcome, period of observation, and duration of study follow-up is not surprising. For example, in the 15 studies included in the present meta-analysis, the period of observation was from 1963 to 2010. The older studies Our study has several strengths. First, our meta-analysis included 15 studies, involving 4,487,662 participants and 17,945 liver cancer cases. The large sample size afforded increased statistical power. We were also able to carry out multiple subgroup analyses and evaluate heterogeneity and the presence of publication bias. Moreover, the present meta-analysis included an approved quality evaluation system. Thus, it minimized potential bias. Further, the likelihood of important selection or publication bias in our meta-analysis is small.Cochrane Handbook for Systematic Reviews of Interventions, Egger's teat typically has low power. Thus, even when they do not provide evidence of funnel plot asymmetry, bias, including publication bias, cannot be excluded. Thus, publication bias remains one possible alternative explanation for our positive finding of an association between gallstones and cholecystectomy and an increased risk of liver cancer.However, several limitations should be acknowledged when interpreting our findings. First, there was significant evidence of considerable heterogeneity among almost all the analyses, despite stratifying the data into subgroups based on several variables. Second, all studies included in our analysis were observational studies, which have inherent limitations. Therefore, confounding factors and bias might have distorted the results In conclusion, although this meta-analysis has some limitations, particularly the heterogeneity of the studies, the results suggest that a history of gallstones and cholecystectomy is associated with an increased risk of liver cancer. This issue would probably be best addressed using well-designed prospective studies with adequate exposure measurements, accurate case definition, and careful adjustment for major confounders.Table S1References for studies excluded from the full text studies review.(DOC)Click here for additional data file.Table S2Newcastle–Ottawa scale for assessment of study quality. A case control studies; B cohort studies.(DOC)Click here for additional data file.Checklist S1PRISMA check list.(DOC)Click here for additional data file.


HIF-1 (hypoxia-inducible factor 1) is a transcriptional activator that functions as a critical regulator of oxygen homeostasis. Recently, a large number of epidemiological studies have investigated the relationship between HIF-1α C1772T/G1790A polymorphisms and cancer susceptibility. However, the results remain inconclusive. Therefore, we performed a meta-analysis on all of the available case-control studies to systematically summarize the possible association.A literature search was performed using PubMed and the Web of Science database to obtain relevant published studies. Pooled odds ratios (ORs) and corresponding 95% confidence intervals (CIs) for the relationship between HIF-1α C1772T/G1790A polymorphisms and cancer susceptibility were calculated using fixed- and random-effects models when appropriate. Heterogeneity tests, sensitivity analyses and publication bias assessments were also performed in our meta-analysis.vs. CC: OR = 1.63, 95% CI = 1.02-2.60; CT + TT vs. CC: OR = 1.15, 95% CI = 1.01-1.34; TT vs. CT + CC: OR = 2.11, 95% CI = 1.32-3.77; T vs. C: OR = 1.21, 95% CI = 1.04-1.41). Similarly, the statistically significant association between the HIF-1α G1790A polymorphism and cancer susceptibility was found to be consistently strong in all of the genetic models. Moreover, increased cancer risk was observed when the data were stratified by cancer type, ethnicity and the source of controls.A total of 40 studies met the inclusion criteria were included in the meta-analysis: 40 studies comprised of 10869 cases and 14289 controls for the HIF-1α C1772T polymorphism and 30 studies comprised of 7117 cases and 10442 controls for the HIF-1α G1790A polymorphism. The results demonstrated that there were significant association between the HIF-1α C1772T polymorphism and cancer susceptibility under four genetic models  contains supplementary material, which is available to authorized users. Human cancer is a major cause of death in the world, and it is estimated that the number of new cases will increase to more than 15 million in the coming decade, creating a substantial worldwide public health burden . Various2 concentration of less than 6%. Under hypoxic condition, HIF-1α ubiquitination decreases dramatically, resulting in an accumulation of the protein, while under normoxic condition, HIF-1α is rapidly degraded through von Hippel-Lindau (VHL)-mediated ubiquitination and proteasomal degradation [Hypoxia-inducible factor 1 (HIF-1) is a transcriptional activator that functions as a critical regulator of oxygen homeostasis. It is a heterodimer composed of two subunits, HIF-1α and HIF-1β, which dimerize and bind to DNA via the basic helix-loop-helix Per/Arnt/Sim (bHLH-PAS) domain , 6. HIF-radation –10. HIF-radation , 12.The HIF-1α gene is located on chromosome 14q21-24, and recent studies have shown that there are a total of 35 common single nucleotide polymorphisms (SNPs) throughout the HIF-1α gene in Caucasian and Asian population –15. Two In recent years, the HIF-1α gene has been a research focus in the scientific community, and many epidemiological studies have been performed to assess the association between HIF-1α C1772T/G1790A polymorphisms and cancer susceptibility. However, the results of the different studies are conflicting. Hence, we performed a meta-analysis of all of the eligible studies to clarify the role of HIF-1α C1772T/G1790A polymorphisms in cancer development.We performed a computerized literature search of the PubMed and Web of Science databases to identify all of the relevant studies of cancer that contained sufficient genotyping data for at least one of the two polymorphisms, HIF-1α C1772T or HIF-1α G1790A. The search strategy was designed by two researchers and included the following keywords: “HIF-1 OR hypoxia-inducible factor-1” and “polymorphism”, and the last search was updated on September 20th, 2013. To obtain all eligible publications, we also manually reviewed the references of the selected articles to identify other potential eligible publications. Articles investigating the association between cancer risk and the HIF-1α polymorphisms were identified with no language restriction.The studies selected were required to meet the following criteria: 1) evaluate the association between the HIF-1α C1772T and/or HIF-1α G1790A polymorphisms and cancer risk; 2) use a human case-control design; 3) contain sufficient published data for the estimation of an odds ratio (OR) with a 95% confidence interval (CI).Data were extracted from all of the eligible publications by two investigators (Yan and Chen) independently, according to the inclusion criteria listed above. Disagreements between the two investigators were resolved by discussion until a consensus was reached. The following information was extracted from each of the included publications: the first author’s name, publication data, country of origin, ethnicities of the sample population , cancer type, source of control group , total number of cases and controls, and the number of cases and controls with the HIF-1α C1772T/G1790A polymorphisms.P < 0.05 was considered to be statistically significant . For HIF-1α C1772T polymorphism, we examined the overall ORs and compared the cancer incidence using the allelic model (T versus C), homozygote model (TT versus CC), heterozygote model (TC versus CC), dominant model (TT + TC versus CC), recessive model (TT versus TC + CC). For HIF-1α G1790A polymorphism, we evaluated the risk in the allelic model (A versus G), homozygote model (AA versus GG), heterozygote comparison model (GA versus GG), dominant models (AA + AG versus GG), and recessive model (AA versus AG + GG). Subgroup analyses were also conducted by ethnicity, cancer type (“other cancer groups” means any cancer types with less than two separate publications) and source of controls. Statistical heterogeneity was estimated by a chi-square based Q-test, and when P < 0.05, the heterogeneity was considered to be significant. We combined all of the values from each individual study using the fixed-effect model and the random-effect model. When P > 0.05, the effects were assumed to be homogenous, and the fixed-effect model (the Mantel-Haenszel method) was used [P < 0.05, the random-effect model (the DerSimonian and Laird method) was more appropriate [2 (I2 = 100% × (Q-df)/Q) was used to quantitatively estimate the heterogeneity, and the percentage of I2 was used to describe the extent of the heterogeneity, I2 < 25%, 25-75% and >75% represent low, moderate and high inconsistency, respectively [P < 0.05 was considered to be statistically significant. Sensitivity analyses were carried out to assess the stability of the results by conducting analysis of studies with controls in HWE. Finally, the Begg’s funnel plot and Egger’s test were utilised to estimate the publication bias [P-values were two-sided and a P of < 0.05 was considered to be statistically significant.The strength of the association between the HIF-1α C1772T/HIF-1α G1790A polymorphisms and cancer risk was measured by ORs with 95% CIs. The statistical significance of the pooled OR was calculated by the Z test, a was used . When P ropriate . The intectively , 24. In et al.) [Through the literature search and selection, a total of 40 eligible studies met the inclusion criteria and were included in our meta-analysis. One study   providedet al.) , 26–63, et al.) , 62, 63.et al.) , 60–62. vs. CC: OR = 1.63, 95% CI = 1.02-2.60; CT + TT vs. CC: OR = 1.15, 95% CI = 1.01-1.34; TT vs. CT + CC: OR = 2.11, 95% CI = 1.32-3.77; T vs. C: OR = 1.21, 95% CI = 1.04-1.41). In the subgroup analysis by cancer type, the HIF-1α C1772T polymorphism significantly increased the risk of breast cancer in Asians , other cancers  and lung cancer . When the data was stratified by ethnicity, the HIF-1α C1772T polymorphism was significantly correlated with an increased cancer risk in Asian population  and Caucasian population . In the analysis stratified by the sources of controls, a significant association was observed in the hospital-based group  and the population-based group . Sensitivity analyses were carried out to assess the stability of the results by conducting analyses of studies with controls in HWE. The results showed significantly increased cancer risk . The other results for the HIF-1α C1772T polymorphism were similar to those when the studies with controls not in HWE were included. The main results of this pooled analysis are shown in Table For the HIF-1α C1772T polymorphism, the overall results from the eligible studies demonstrated a significant association between the HIF-1α C1772T polymorphism and an increased cancer risk in four genetic models . In the subgroup analysis by cancer type, a significant association was observed in lung cancer , pancreatic cancer  and renal cancer . When the data were stratified by ethnicity, significantly increased cancer risk was observed in Asian population and Caucasian population. When the studies were stratified by the source of controls, a significant association was observed for population-based controls under the homozygote model, the dominant comparison model and the allelic model. Sensitivity analyses were conducted after the removal of the studies with controls not in HWE, the results for the HIF-1α G1790A polymorphism were similar to those when the studies with controls not in HWE were included. Table For HIF-1α G1790A polymorphism, as shown in Table There was significant heterogeneity observed in the allelic comparison model, the dominant comparison model and the heterozygote comparison model Tables  and 3, aWe performed sensitivity analysis by removing each individual study (including the restudies with controls not in HWE) sequentially for both the HIF-1α C1772T and the HIF-1α G1790A polymorphism . For the HIF-1α G1790A polymorphism, no visual publication bias was detected in the funnel plot .A Begg’s funnel plot and Egger’s test were used to assess the publication bias in this meta-analysis. As shown in Figure 2-regulated subunit that determines the function of HIF-1. HIF-1α upregulates the expression of genes whose protein products function to increase O2 availability or to allow metabolic adaptation to O2 deprivation, including VEGF, Epo, NOS2 and others. Most of these aforementioned proteins have been implicated in tumor development and progression [http://www.ncbi.nlm.nih.gov/SNP). Two polymorphisms, C1772T (rs11549465) and G1790A (rs11549467), result in an amino acid substitution of proline to serine and alanine to threonine, respectively, and the present studies show that C1772T (rs11549465) is not in substantial linkage disequilibrium (LD) with G1790A (rs11549467) (R2 = 0.002). Under normoxic condition, the hydroxylation of proline 402 and proline 564 occurs within the oxygen-dependent degradation (ODD) domain of HIF-1α, and HIF-1α is rapidly degraded. The two SNPs examined here are located within the ODD/pVHL binding domain in exon 12 of the HIF-1α gene and may enhance the transcription activity of the HIF-1α gene by causing structural changes, increasing the stability of HIF-1α protein and affecting the expression of downstream target genes [et al. in 2009 [et al. reported that the HIF-1α C1772T polymorphism correlates with urinary cancer risk in Caucasian population, and the G1790A polymorphism may increase the risk of prostate cancer [HIF-1 is a heterodimeric transcription factor and a key regulator of the cellular response to hypoxia . It is cgression , 64, 65.gression , 35, 66.et genes , 14, 17. rs115494, result e cancer . Due to et al.[Regarding the HIF-1α C1772T polymorphism, our results suggested a significant association in four genetic comparison models, providing convincing evidence that the HIF-1α C1772T polymorphism may be a risk factor in cancer development. When sensitivity analyses were performed, the results were similar to those when the studies with controls not in HWE were included, suggesting that our results were very robust. Moreover, when the data were stratified by cancer type, a significant association was observed between the C1772T polymorphism and breast cancer in Asians. This may be due to the specific genetic variant induced over-expression of HIF-1 under hypoxic condition in breast cancer cells, and the different life style, ethnicity and body composition between Asians and Caucasians, which could contribute to the results. A significant association was also observed in lung cancer. When subgroup analyses were performed according to ethnicity and source of controls, a significant association was found in Asian population, Caucasian population and in hospital-based studies. Zhao et al. reportedet al. indicated that even though they did not found any significant differences in genotype for G1790A between lung cancer patients and healthy controls, however, the G1790A variant allele was significantly higher in lung cancer patients, and TP53 LOH and 1p34 LOH were more frequently observed in individuals with the HIF-1α G1790A polymorphism, suggesting that this polymorphism may induce mutations in some tumor suppressor genes involved in lung cancer development [For the HIF-1α G1790A polymorphism, the pooled results from all of the eligible studies suggested that the G1790A polymorphism in HIF-1α is significantly associated with an increased cancer risk in all of the genetic models. We also conducted subgroup analyses based on the cancer type, ethnicities and source of controls. In the subgroup analysis according to cancer type, the results suggested that the HIF-1α G1790A polymorphism significantly increased the risk of lung cancer, renal cancer, oral cancer and pancreatic cancer, but the CI for the oral cancer subgroup was very wide. This may be due to the lack of mutant homozygotes detected, and the association could have been caused by chance. More studies based on large populations should be prusued. The study reported by Putra elopment . Here, wTo a certain extent, our meta-analysis still includes several limitations that should be interpreted and taken into consideration. First, in the era of GWAS, researchers can obtain the GWAS data for these two SNPs from all cancer studies and conduct a meta-analysis with the GWAS data instead of relying on published data, which may be biased toward positive findings. Second, the lack of observations concerning gene-gene and gene-environment interactions could influence our results. Third, although the total number of studies was not small, there were still not sufficient eligible studies for us to analyze different types of cancers, such as colorectal carcinoma, renal cell carcinoma or glioma; more studies are needed to explore the potential relationship between HIF-1αC1772T/G1790A polymorphisms and cancer susceptibility. Forth, the lack of detailed original data, such as the age and sex of the populations, smoking status, or alcohol consumption in the eligible studies may influence our extended analyses. However, our meta-analysis also has many advantages. First, we searched all possible publications, and the total number of eligible studies was much larger than other previously published meta-analyses; therefore, our results are more convincing. Second, no publication bias was detected in our meta-analysis. Finally, all of the data were extracted from well-selected studies, providing stronger statistical power for our study.In conclusion, this meta-analysis provides powerful evidence that both the C1772T and G1790A polymorphisms in the HIF-1α gene may contribute to individual susceptibility to cancers. It will be necessary to perform additional research to evaluate the relationship between HIF-1α C1772T/G1790A polymorphisms and cancer risk. Moreover, large sample case-control studies assessing gene-to-gene and gene-to-environment interactions are required to verify these findings.Qing Yan, Pin Chen and Songtao Wang are joint first authors.The influence of individual studies on the summary odds ratio (OR) for the HIF-1α C1772T polymorphism.(TIFF 271 KB)Additional file 1:


The EGFR R521K polymorphism has been shown to reduce the activity of EGFR; however, the association between EGFR R521K polymorphism and the risk of cancer remains inconclusive; therefore we performed a meta-analysis to evaluate the relationship between EGFR R521K polymorphism and susceptibility to cancer. Our results suggest that the EGFR R521K polymorphism is not associated with risk of cancer, but the different chemosensitivity to anticancer drugs may need further investigation. Cancer is one of the leading causes of death in humans worldwide. It is not a single, well-defined, but a generalized term of heterogeneous diseases including more than a hundred types. Although each type presents an independent feature, the essential pathological characteristics are referred to as aberrant cell proliferation, invasion, and migration.Many studies have investigated the role of human epidermal growth factor receptor  in cancer. Abnormal expression or activity of EGFR observed in many types of cancer, like breast cancer, colorectal cancer, non-small-cell lung cancer, head and neck squamous cell carcinoma, has been considered as an important marker in tumorigenesis, which resulted in more aggressive tumor phenotype, higher recurrence rate, and poorer prognosis .α transforming growth factor (TGF-α). The binding of ligand triggers the activation of MAPK, PI3K/Akt/mTOR, and JAK/STAT downstream signalling pathway, which are responsible for regulation of numerous tumorigenic functions such as cancer cell proliferation, antiapoptosis, and metastasis  to adenine [A] mutation leading to an arginine [R] to lysine [K] substitution at codon 521 located in the CR2 domain and resulted in an attenuated activity of the receptor . SeveralA comprehensive search in PubMed, Web of Science, OVID, and Cochrane Library regarding R521K polymorphism was performed. The following search words were used: “EGFR,” “ErbB1,” “HER1,” “polymorphism,” “R497K,” “R521K,” “rs11543848,” “rs2227983,” “142285G>A,” and “cancer.” We also traced the references of the articles and reviews to include the potentially eligible original reports.All articles were included in the meta-analysis if they were (1) case-control studies, (2) studies on the relationship of R521K polymorphism and cancer risk, (3) sufficient data for the frequencies of alleles and genotypes in cases and controls.Articles were excluded if they were (1) not related to cancer, (2) of insufficient information, (3) animal, bacteria, or cell line studies, (4) studies on the basis of family or twins rather than random general ones, and (5) repeatedly published data.Two investigators extracted the information independently including first author, journal, year of publication, country of origin, cancer type, genotyping method, sample size, source of control groups, frequency of genotypes and alleles in cases and controls, and the Hardy-Weinberg equilibrium (HWE) of genotype distribution in controls. Any disagreements between the two investigators were resolved by consensus.http://ihg.gsf.de/cgi-bin/hw/hwa1.pl). The odds ratios (ORs) and the corresponding 95% confidence intervals (CIs) were used to assess the association between the R521K polymorphism and cancer risk. The pooled ORs were performed in allelic genetic model (A versus G), codominant model (AA versus GG and GA versus GG), recessive genetic model (AA versus GA + GG), and dominant genetic model (AA + GA versus GG). Subgroup analyses were also conducted based on ethnicity and cancer type. Heterogeneity between studies was checked by Chi-square based Q statistics, and the I2 index which expresses the percentage of the total variation across studies was also tested. If the P value was <0.05 or the I2 index was ≥50%, the random-effect model was adopted; otherwise the fixed-effect model was adopted [HWE that has not been reported in the literatures was assessed with the De Finetti program . In subgroup analysis, only a significant association between R521K polymorphism and reduced cancer risk was observed in the allelic genetic model in gastric cancer, but not in other genetic models . No significant difference in other types of cancer and ethnicities was found.The summary of the results for the association between EGFR R521K polymorphism and cancer susceptibility was shown in There was only low between-study heterogeneity in homozygote model (AA versus GG) and recessive genetic model (AA versus GA + GG) both in overall and subgroup analyses, but high heterogeneity was observed in allelic genetic model (A versus G), heterozygote model (GA versus GG), and dominant genetic model (AA + GA versus GG). Therefore the sensitivity analysis and metaregression analysis were performed to clarify the source of heterogeneity. In the sensitivity analysis, a single study was omitted each time to investigate the influence of the individual data on the pooled ORs. No significant alteration occurred in this procedure. In the metaregression analysis, neither cancer type nor ethnicity could explain the significant between-study heterogeneity.P = 0.285; AA versus GG: P = 0.691; AA + GA versus GG: P = 0.713; AA versus GA + GG: P = 0.419; and GA versus GG: P = 0.319).The results of the Begg funnel plots indicated no evidence of obvious asymmetry. The results of Egger's test also suggested little evidence of publication bias in the overall analysis  family, whose members share a highly conserved extracellular ligand binding domain, a single-helix transmembrane segment, a tyrosine kinase juxtamembrane segment, and a regulatory carboxyterminal tail. The extracellular region of EGFR is composed of four domains: Domains I (L1), II (S1 or CR1), III (L2), and IV (S2 or CR2). L1 and L2 are members of the leucine-rich repeat family and are responsible for ligand binding. CR1 and CR2 are homologous cysteine-rich domains. In the absence of ligand, a ~25 Å ing site , 28. Theing site .α, a ligand which is related to cell proliferation, differentiation, and invasion after binding to EGFR. However, another study demonstrated that the CR1/CR2 tether structure exerted only a limited effect on EGF binding, EGFR activation, and cell signalling [n repeat polymorphism is associated with a better survival rate in cancer [The R521K polymorphism is located in the CR2 domain and presents a lower affinity to TGF-gnalling . Take ougnalling . Neverthn cancer –36, whiln cancer . Furthern cancer , 38–41 bn cancer , 42, 43.n cancer , 45.Studying heterogeneity was a considerable problem in our meta-analysis. The heterogeneity existed in allelic genetic model (A versus G), heterozygote model (GA versus GG), and dominant genetic model (AA + GA versus GG) in overall and subgroup analysis. The metaregression analysis did not identify the source of heterogeneity. In sensitivity analysis, heterogeneity reduced after one literature was excluded, but the pooled ORs did not alter remarkably, implying that the results were statistically stable and reliable. The Begg funnel plot and Egger's test were also negative for publication bias.There were some limitations in our meta-analysis. First, our meta-analysis was based on only 13 studies including a small sample size; this may lead to a relatively low statistical power, especially in subgroup analysis. Second, not all the sources of controls were population based; this may induce deviations from the overall population. Third, no gene-environment or gene-gene interactions were considered which may be involved in the susceptibility to cancer. Fourth, the relation between R521K polymorphism and the chemosensitivity of anticancer drugs remains to be elucidated.In conclusion, there is no significant association between EGFR R521K polymorphism and risk of cancer. However, further studies based on larger, stratified case-control populations and the effect of gene-gene, gene-environment, and the chemosensitivity of anticancer drugs are still needed to be evaluated.


The polymorphisms in the three main heat shock protein 70  genes were identified to be associated with cancer risk. However, the results are inconsistent. We perform a meta-analysis to evaluate the association between the three HSP70 polymorphisms and cancer risk. Relevant studies were identified using PubMed, Web of Science, Chinese National Knowledge Infrastructure (CNKI), and Wanfang databases up to March 29, 2014. The cancer risk associated with the HSP70 polymorphisms was estimated for each study by odds ratios (OR) together with its 95% confidence interval (CI), respectively. Twenty case-control studies from eighteen publications were included; a significant association was observed for HSP70-2 polymorphism , while there was no significant association for HSP70-1 and HSP70-hom polymorphisms. Besides, in stratification analyses by ethnicity, cancer type, and source of control, significant association was detected for HSP70-2 polymorphism, while for HSP70-hom polymorphism, we found a significant association in hospital-based population under homozygote comparison model. This meta-analysis suggests that the HSP70-2 polymorphism rather than HSP70-hom and HSP70-1 polymorphisms was associated with the risk of cancer. Cancer is recognized as one of the leading causes of death in economically developed countries as well as in developing countries. According to the estimation of GLOBOCAN, approximately 12.7 million new cases and 7.6 million deaths of cancer had occurred in 2008, it has become a major public health challenge .Heat shock proteins (HSPs) are evolutionarily highly conserved stress proteins expressed and induced by heat shock, infection, inflammation, ischemia hypoxia, oxidative stress, carcinogens, and so on , 5. MammTo date, several studies have investigated the association between the three HSP70 polymorphisms and risk of cancer –31. HoweWe searched for relevant studies before March 29, 2014, by using electronic PubMed, Web of Science, Chinese National Knowledge Infrastructure (CNKI), and Wanfang databases with the following terms: “heat shock protein 70 or HSP70,” “genetic polymorphism or polymorphisms or variant or SNP,” and “cancer or carcinoma or tumor.” The search was restricted to humans and without language restrictions. Additional studies were identified by a hand search of references of original or review articles on this topic. If more than one geographic or ethnic heterogeneous group was reported in one report, each was extracted separately.The inclusion criteria were as follows: (1) studies that evaluated the association between the HSP70 polymorphisms and cancer, (2) a case-control study design, and (3) studies that had detailed genotype frequency of cases and controls or could be calculated from the paper text, while major exclusion criteria were (1) case-only study, case reports, and review articles, (2) studies without the raw data of the HSP70 genotype, and (3) studies that compared the HSP70 variants in precancerous lesions.P value). All data were extracted by two investigators independently, using the same standard. The results were compared and disagreements were resolved by consensus.The following information was extracted from each eligible publication: the first author's name, year of publication, country of origin, ethnicity, cancer type, source of control, genotyping methods, number of cases and controls, and Hardy-Weinberg equilibrium (HWE) in controls (χ2-test-based Q statistic test was performed to assess the between-study heterogeneity [I2 test. When a significant Q test (P < 0.05) or I2 > 50% indicated heterogeneity across studies, the random effects model was used [χ2 test. We performed stratification analyses on ethnicity, cancer type, and source of controls. Analysis of sensitivity was performed to evaluate the stability of the results. Finally, potential publication bias was investigated using Begg' funnel plot and Egger's regression test [P < 0.05 was considered statistically significant.The risk of cancer associated with the HSP70 polymorphisms was estimated for each study by odds ratio (OR) and 95% confidence interval (95% CI). Four different ORs were calculated: dominant model (the combined variant homozygote and heterozygote versus the wild-type homozygote), recessive model (the variant homozygote versus the combined heterozygote and wild-type homozygote), heterozygote comparison (heterozygote versus the wild-type homozygote), and homozygote comparison (variant homozygote versus the wild-type homozygote). A ogeneity . We alsowas used  or else was used . In addiion test , 36. P <All analyses were performed by the Cochrane Collaboration RevMan 5.2 and STATA package version 12.0 .After an initial search, a total of 132 published articles relevant to the topic were identified. According to the inclusion criteria, 21 studies –31 with For HSP70-2 polymorphism, 17 studies with 2134 cases and 2818 controls were identified. Overall, a significant association was found  . In straFor HSP70-hom polymorphism, 10 studies with 2042 cases and 2661 controls were identified. Overall, no significant association was found under all genetic models . When the analysis was stratified by ethnicity, similar results were observed among Asian, European, African, and mixed populations. While, in stratified analysis by source of controls, a significant association was found in the hospital-based subgroup  but not in the population-based subgroup .For HSP70-1 polymorphism, 5 studies with 1558 cases and 2075 controls were identified. The pooled results suggested that no significant association was found in overall analysis  .I2 = 81%, P < 0.00001, recessive model: I2 = 87%, P < 0.00001, AG versus AA: I2 = 74%, P < 0.00001, GG versus AA: I2 = 88%, P < 0.00001; HSP70-hom: dominant model: I2 = 87%, P < 0.00001; recessive model: I2 = 60%, P = 0.008; TC versus TT: I2 = 83%, P < 0.00001; CC versus TT: I2 = 68%, P = 0.001; HSP70-1: dominant model: I2 = 89%, P < 0.00001; recessive model: I2 = 79%, P = 0.0008; GC versus GG: I2 = 86%, P < 0.0001; CC versus GG: I2 = 83%, P = 0.0001)  . TherefoP = 0.039, HSP70-hom polymorphism: P = 0.537, and HSP70-1 polymorphism: P = 0.367).Begg's funnel plot and Egger's test were performed to assess the potential publication bias in the available literature. The shape of funnel plots did not reveal any evidence of funnel plot asymmetry (data not shown). Egger's test also showed that there was no statistical significance for the evaluation of publication bias under dominant model  for the date analysis may cause serious confounding bias.In summary, this meta-analysis suggested that the HSP70-2 polymorphism rather than HSP70-hom and HSP70-1 polymorphisms was associated with the risk of cancer. However, large and well-designed studies taking into consideration gene-gene and gene-environment interactions are warranted to validate our findings.


The 2007 World Cancer Research Fund/American Institute for Cancer Research expert report concluded that foods containing vitamin C probably protect against esophageal cancer and fruits probably protect against gastric cancer. Most of the previous evidence was from case–control studies, which may be affected by recall and selection biases. More recently, several cohort studies have examined these associations. We conducted a systematic literature review of prospective studies on citrus fruits intake and risk of esophageal and gastric cancers.PubMed was searched for studies published until 1 March 2016. We calculated summary relative risks and 95 % confidence intervals (95 % CI) using random-effects models.With each 100 g/day increase of citrus fruits intake, a marginally significant decreased risk of esophageal cancer was observed . The associations were similar for squamous cell carcinoma  and esophageal adenocarcinoma . For gastric cancer, the nonsignificant inverse association was observed for gastric cardia cancer , but not for gastric non-cardia cancer . Consistent summary inverse associations were observed when comparing the highest with lowest intake, with statistically significant associations for esophageal  and gastric cardia cancers .Citrus fruits may decrease the risk of esophageal and gastric cardia cancers, but further studies are needed.The online version of this article (doi:10.1007/s10552-016-0755-0) contains supplementary material, which is available to authorized users. Esophageal and gastric cancers are the eight and the fifth most common cancers worldwide, respectively . EsophagHelicobacter Pylori infection. Despite the possibility of preventing non-cardia gastric cancer by treating H. Pylori infection, there are concerns with possible adverse consequences of the antibiotic treatment, such as development of antibiotic resistance and alterations of the intestinal microbiota [Tobacco use is a risk factor for esophageal and gastric cancers. Alcohol and tobacco use are the main risk factors for esophageal SCC . Due to crobiota . There iDiet may also play a role on the development of esophageal and gastric cancers. In 2007, the World Cancer Research Fund/American Institute for Cancer Research (WCRF/AICR) Second Expert Report concluded that there was evidence that high total intake of salt probably increases the risk of gastric cancer, and that vegetables and fruits intake probably protects against esophageal and gastric cancers . With reCitrus fruits are rich in vitamin C, and foods containing vitamin C were judged probably to protect against esophageal cancer in the WCRF/AICR Second Expert Report . Much ofA recent meta-analysis of cohort studies reported nonsignificant inverse association between citrus fruits intake and the risk of gastric cancer for the comparison of the highest versus the lowest intakes . Howeverhttp://www.wcrf.org/int/research-we-fund/continuous-update-project-cup. In addition, reference lists of relevant reviews identified in the search and of the studies included in the meta-analysis were screened for any further publications.All cohort studies identified in the systematic literature review for the WCRF/AICR Second Expert Report  were indThe following inclusion criteria were applied for studies included in this meta-analysis: (a) cohort, nested case–control or case-cohort design; (b) reported estimates of the relative risk  with confidence intervals (CI); (c) reported quantifiable measure of citrus fruits intake. If several publications using the same study population were identified, the one with the largest number of cases was selected.The following data were extracted from each study: the first author’s last name, publication year, country in which the study was conducted, study name, follow-up period, sample size, sex, age, number of cases, dietary assessment method , exposure, frequency or amount of intake, associated RR and corresponding 95 % CI, and adjustment variables. The search and data extraction for the systematic literature reviews of esophageal cancer and gastric cancer prior to January 2006 was conducted by the WCRF/AICR Second Expert Report teams at the Pennsylvania State University and the University of Leeds, respectively . The seaWe conducted dose–response meta-analyses and summarized the associations for the highest compared to the lowest citrus fruits intake reported in the studies using random-effects models .http://www.wcrf.org/sites/default/files/protocol_oesophageal_cancer.pdf). The dose–response was expressed for an increment of 100 g/day of citrus fruits. We used the multivariable adjusted RR from each study. The EPIC study [When not provided in the publications, the linear dose–response trends were derived from the natural logs of the RRs and CIs across categories of citrus fruits intake, using the method by Greenland and Longnecker . For thiIC study , 19 repo2 statistics, with low and high heterogeneity extent indicated by I2 values below 30 % or substantially higher than 50 % [We first estimated summary RR for all esophageal and gastric cancers, respectively. For these analyses, the RRs for men and women were combined using fixed-effect meta-analysis before pooling. When RRs were reported by cancer subtypes only, we estimated the combined RR of gastric cardia and non-cardia or esophageal adenocarcinoma and squamous cell carcinoma using Hamling’s method . The methan 50 % .Subgroup analyses were conducted to assess possible sources of heterogeneity, as well as study quality. The predefined factors to explore were sex, outcome type, geographic location, duration of follow-up, number of cases, publication year, and adjustment for confounders including smoking, alcohol intake and adiposity (as measured by BMI), when the number of studies allowed it.Publication bias was assessed with Egger’s test  and visuFlowcharts of the search are provided as an online resource Fig. a, b. SevMain study characteristics are shown in Table All measures of association included in the meta-analyses were adjusted for multiple confounding factors, albeit defined differently in the studies, including alcohol , 25–27, Gastric or esophageal cancers were primary outcomes in all but two studies , 25 thatFive studies were conducted in Asia , 27, 28,I2 = 0 %, Pheterogeneity = 0.83) . The summary RR for the highest compared with the lowest intake was 0.77 (95 % CI 0.64–0.91), with low heterogeneity   Fig. a. There I2 = 0 %, Pheterogeneity = 0.58) and 0.87  for SCC with low heterogeneity  , no differences emerged across study characteristics, including adjustment factors Table . There iI2 = 31 %, Pheterogeneity = 0.34)       –19, 26. p = 0.25).Summary risk estimates observed in subgroup analyses for all gastric cancers were mostly similar to that in the overall analysis, with exceptions in some subgroups where a positive association was observed. Estimates of risk were below 1 in studies adjusted for smoking and alcohol and BMI but not in the unadjusted studies Table . Signifip for interaction =0.07). Other studies in the review explored the interaction of smoking and intake of total fruits and vegetables, or fruits. In general, no significant interactions with smoking were observed. In the NIH-AARP study, the risk estimates of adenocarcinoma and SCC for total fruits and vegetable intakes appeared similar in smokers, non-smokers, and current smokers [p for interaction = 0.25; 0.15; and 0.49, respectively) [p for interaction = 0.27) [One study reported on the interaction of smoking status and citrus fruits intake in relation to esophageal or gastric cancers. In the EPIC study , the inv smokers . In the  smokers . In the ctively) . In a Ch = 0.27) .In these meta-analyses of cohort studies, citrus fruits intake was marginally associated with reduced risks of esophageal and gastric cardia cancers. No association with non-cardia gastric cancers was observed. Similar results were observed for adenocarcinomas and SCC of esophagus.Citrus fruits are rich in vitamin C that could influence cancer risk by scavenging reactive oxygen species, protecting mucosal tissues from the damaging effects of oxidative stress, and inhibiting nitrosamine formation in the stomach . The resIn addition to high vitamin C content, citrus fruits contain a wide range of bioactive compounds such as citrus flavonoids, carotenoids, and limonoids. Experimental studies have demonstrated that these bioactive components may protect DNA, regulate cell growth, and induce apoptosis –38.The main limitation of this meta-analysis is the small number and limited power of published studies on citrus fruits intake, esophageal and gastric cancer risks, and the unexplained heterogeneity of the inverse association of citrus fruits intake and gastric cancer cardia in the three studies identified , 19, 26.p for interaction =0.07). However, other studies included in the meta-analysis [p for interaction >0.15) [The observed inverse associations could be due to residual confounding by smoking. In the EPIC study , the invanalysis , 19, 26 analysis  was simianalysis , there wn >0.15) , 40, havn >0.15) , and mayn >0.15) .Measurement error of diet may have attenuated the risk estimates. Only the EPIC cohort corrected for dietary measurement error , 19. WheIn conclusion, there is evidence from cohort studies that citrus fruits may decrease the risk of esophageal and cardia gastric cancers, but the data are still limited.Supplementary material 1 (DOCX 40 kb)Below is the link to the electronic supplementary material.


Much attention has been directed to the association between cancer risk and rs2066827 polymorphism of the CDKN1B gene. However, the results are indefinitive and inconclusive. This study was devised to evaluate the hypothesis that rs2066827 polymorphism is associated with the risk of cancer.Ph > 0.05), and random-effects pooled ORs were estimated by the DerSimonian–Laird method (Ph < 0.05).Computer-based databases  were used to seek all case–control studies evaluating rs2066827 polymorphism and susceptibility to cancer. The genetic risk was assessed by calculating pooled odds ratio (OR) with its corresponding 95% confidence interval (CI). Fixed-effects pooled ORs were calculated by the Mantel–Haenszel method . We observed a moderately decreased risk of ovarian cancer based on 1829 cases and 2868 controls . A slightly deceased risk of cancer was also indicated in Caucasians consisting of 6707 cases and 8279 controls .These data suggest that carriage of a TG genotype at rs2066827 polymorphism may be associated with decreased susceptibility to cancer, ovarian cancer in particular. Uncontrolled cell proliferation and damage to genes involved in regulating the cell cycle are known causes of human carcinogenesis.3 Cyclin E forms a complex with CDK2. Cyclin E/CDK2 genes are master mediators of the cell cycle progression in the G1 phase and of the G1-S phase transition. In cancerous tissues and cancer cell lines, cyclin E can be proteolytically separated into low molecular weight (LMW) isoforms which are biologically functional. Overexpressed LMW increases cell proliferation rate and deregulates the cell cycle of mammary epithelial cells.4Cyclin-dependent kinases (CDKs) and CDK inhibitors (CKIs) are important regulators of cell cycle progression.CDKN1B) gene mapped to chromosome 12p13. The cell cycle inhibitor protein hampers enzymatic activation of cyclin E/CDK2 complexes and induces cell cycle arrest at G1 phase.5 Lower expression of cyclin-dependent kinase inhibitor CDKN1B related to tumorigenesis and advanced clinical stage has been identified in multiple types of cancer, such as cancers of prostate, gastric, laryngeal, colorectal, and breast.10p27 protein is encoded by the cyclin-dependent kinase inhibitor 1B (CDKN1B gene have been linked to reduced transcription rate and lower protein CDKN1B levels.11 There lies an SNP (T to G) at codon 109 that causes an amino acid change from valine to glycine .12 The common SNP may affect the normal function of CDKN1B by altering its transcription efficiency and protein levels, leading to cell cycle dysregulation and cancer cell behavior.13 A series of epidemiological studies have demonstrated increased genetic susceptibility to cancer ascribed to rs2066827 polymorphism.15 However, several recent studies failed to replicate this finding. We therefore performed a meta-analysis incorporating 17 studies with 9038 cancer cases and 11,596 controls to determine the association between cancer risk and rs2066827 polymorphism.Single nucleotide polymorphisms (SNPs) of the CDKN1B,” “rs2066827,” and their synonyms  in combination, without language restrictions. Reference lists of all relevant review articles, meta-analyses, and human case–control studies addressing the association of interest were examined to obtain additional usable data. The study was supported by the Research Ethics Committee of Shandong Provincial Hospital Affiliated to Shandong University.Human case–control studies estimating the risk of cancer associated with rs2066827 polymorphism were identified by searching EMBASE, PubMed, and CNKI databases. We used the following keywords: “cancer,” “polymorphism,” “Genetic association studies satisfying all of the following criteria were included in this meta-analysis: a case–control study published in a peer-reviewed journal, examined the association of rs2066827 polymorphism with cancer, reported the genotype frequencies of rs2066827 polymorphism in cases and controls in detail, and genotype frequencies in controls must be in agreement with Hardy–Weinberg equilibrium (HWE). The studies were excluded if published as an abstract with insufficient genotype data, designed as a case-case or case-only study, and comments, editorials, letters, or systematic reviews. When several articles were conducted based on the same patients, the more informative article with a larger sample size was selected for further analysis.Data were extracted on first author, year of publication, country of study, ethnicity of each study population, selection and characteristics of cases and controls, genotyped cases and controls, cancer type, method used for genotyping, frequencies of genotypes, and results of HWE test. Data extraction was done separately by 2 of the investigators. Discrepancies were resolved by consensus involving a senior investigator.2-based Q-test, with P < 0.05 being considered significant. The Mantel–Haenszel method was used to calculate fixed-effects pooled ORs,16 and the DerSimonian–Laird method was chosen to calculate random-effects pooled ORs.17 Subgroup analyses according to ethnicity  and cancer type  were preformed to detect the potential association for each subgroup.Odds ratios (ORs) with the corresponding 95% confidence intervals (CIs) were calculated to assess cancer risk. Pooled ORs were estimated for 3 genetic models  using the fixed-effects or the random-effects model. Heterogeneity across the studies was measured by the χ2 test. Publication bias was estimated by inspection of funnel plot asymmetry and by calculation of Egger's linear regression method intercept.18 Statistical data were analyzed using STATA software package (v.12.0). The significance threshold was set at P < 0.05.Sensitivity analyses were performed to evaluate influence of the individual studies on the pooled ORs. Deviation from HWE in the controls was checked by the goodness-of-fit χ35 As shown in Table The flow diagram of study exclusion and inclusion with specific reasons is shown in Figure There was a significant association observed between rs2066827 polymorphism and cancer risk Table . We finaP < 0.05, Table 24 and Francisco et al29, respectively, were identified as outliers. The results in GG + TG versus TT, and G versus T models were highly homogeneous when the outliers were excluded (available on request). But we noted that the exclusion resulted in significant alternations in the pooled ORs (available on request).We detected significant between-study heterogeneity in the analyses using GG + TG vs. TT, and G vs. T models  . Another possible explanation is the wide disparity of total number .In our meta-analysis of 9038 cancer cases and 11,596 controls participating in 17 studies, we demonstrated a weak protective effect on cancer in relation to the TG genotype. We also observed a moderate protective effect on ovarian cancer and a slight protective effect in Caucasians after stratification by cancer type and ethnicity, respectively. The most likely reason for absence of an association in Asians may relate to the significant difference in minor allele frequency (G). The frequency in the Asian populations was 3.4%, which was significantly lower compared to Caucasians . In addition, we found a novel association for ovarian cancer and Caucasians after excluding the overlapped data.In a previous meta-analysis where a total of 8 studies with 3591 cases and 3799 controls were included, Wei et al43 However, the current sample was too small to detect the possible effects for multiple cancer types and did not allow further stratification analyses by ethnicity for ovarian cancer and breast cancer. Thus, sample insufficiency is one of the limitations. Second, significant heterogeneity presented in some analyses. Exclusion of the identified outliers caused obvious changes in the risk estimates, suggesting that heterogeneity should be considered in interpreting some results. Finally, effects of common confounding factors, including age, gender, tobacco smoking and alcohol consumption, were not assessed in the present study because of data insufficiency.There are some strengths and limitations in our study. An important strength is the markedly expanded sample size which helped to identify some findings not suggested in the previous meta-analysis. Further, we included all studies with usable data without considering language. The exclusion of papers for linguistic reasons may lead to biased results.In summary, our meta-analysis demonstrated some evidence that the rs2066827 polymorphism of the CDKN1B gene may protect against the development of human cancer, especially ovarian cancer and in Caucasians. Additional work with a larger number is recommended to better understand the potential functional role of this nonsynonymous polymorphism in distinct ethnic populations and various cancers.


P = 0.009). Multi-regression analysis showed that TNF-α-308A was also a risk factor for distant tumour metastasis after adjustment for tumour size and lymph node metastasis status . These findings indicate that TNF-α might play a distinct role in the progression of TNBC, especially in distant tumour metastasis of TNBC.Tumour necrosis factor-α (TNF-α) is critical in the regulation of inflammation and tumour progression. TNF-α-308G > A is associated with constitutively elevated TNF-α expression. The purpose of this study was to assess the association between TNF-α-308G > A and breast cancer (BC) risk by subtype and the connection between genotypes and clinical features of BC. A total of 768 patients and 565 controls were enrolled in this study, and genotypes were detected using the TaqMan assay. No effect on susceptibility for any BC subtype was found for the TNF-α-308 polymorphism in our study or in the pooled meta-analysis. This polymorphism was shown to be associated with age at menarche in all BC and in progesterone receptor-negative BC. Interestingly, triple negative breast cancer (TNBC) patients with TNF-α-308A had an increased risk of distant tumour metastasis (OR = 3.80, 95% CI: 1.31–11.02,  The current case-control study investigated the role of TNF-α-308G > A in BC susceptibility by IHC subtype and ER, PR and Her2 status and the relationship between genotypes and clinicopathological characteristics of BC.Because TNF-α is tumourigenic P = 0.275). Of the 768 BC cases, 163 were TNBC and 82 were Her2+, 183 were luminal A, 340 were luminal B, 523 were ER+, 245 were ER−, 585 were PR−, and 686 were Her2− BCs. The genotype frequency of TNF-α-308G > A in the controls was in concordance with Hardy-Weinberg equilibrium (HWE) (P = 0.940).Because the cases and controls were frequency-matched for age, there were no significant differences in the distributions of age between the cases and controls  in the dominant genetic model. According to IHC classification, no association was found between the TNF-α-308 polymorphism and the TNBC , Her2+ , luminal A  or luminal B  subtypes. Consistent with the results for all BCs, no association was observed between the TNF-α-308 polymorphism and ER+ , ER− , PR−  or Her2−  BCs.The allelic frequency of TNF-α-308A was 0.074 in the controls and 0.065 in all BC cases.  = 1.06, 95% CI: 0.88–1.27, P = 0.555; I2 = 80.7%, Phet < 0.001), Asians  or all subjects  (PBegg’s = 0.596). Additionally, the Egger’s test found no significant publication bias (P = 0.375).Due to a lack of information, we conducted an updated meta-analysis to test whether the TNF-α-308 polymorphism is associated with overall BC risk rather than specifically testing its association with the various subtypes. Eighteen studies (including the present study) were included in this analysis, with a total of 13567 BC cases and 15087 controls. The detailed characteristics of each study in the pooled analysis are summarized in 232425262728293031323334363738< 0.001) . Begg’s  = 0.57, 95% CI: 0.35–0.95, P = 0.029). In the TNBC group, the GA and AA genotypes were associated with distant tumour metastasis . After adjusting for tumour size and lymph node metastasis status, this association remained , and in our study, this association was also statistically significant after adjusting for all clinical characteristics .We analysed the association between TNF-α-308G > A and clinical characteristics  by BC subtypes classified using IHC . In all − patients .We analysed the association between TNF-α-308G > A and clinical characteristics in patients with different ER, PR, and Her2 statuses (Supplementary Table 1). A similar association between the GA and AA genotypes and age at menarche to that observed in all BCs was seen in PR− BCs. Interestingly, patients with TNBC and the TNF-α-308A allele had an increased risk of distant tumour metastasis. Our study highlights the effect of the TNF-α gene polymorphism on the progression of BC by subtype. These findings suggest a potential connection between constitutively higher TNF-α expression and the pathogenesis of TNBC.Inflammatory cytokines play critical roles at different stages of tumour development and progression, including invasion and metastasis. Although several studies have reported an association between the TNF-α-308G > A polymorphism and BC risk, the results of those study were inconsistent. Moreover, little is known about this association in the Chinese. In the present study, we found that the TNF-α-308G > A polymorphism was not associated with the risk of BC in the different IHC subtypes or BCs classified based on ER, PR or Her2 status in Chinese, Asians or Caucasians. We provided evidence that the TNF-α-308 polymorphism is associated with age at menarche in all BCs and in PRTNF-α is mainly produced by macrophages and is also expressed in a wide variety of cells, including mast cells, lymphoid cells, endothelial cells, cardiac myocytes, and fibroblastset al. included one study that compared the frequencies of the different TNF-α-308 polymorphism genotypes in patients with benign breast disease and controlsTwo recent meta-analyses reported that the TNF-α-308GA and AA genotypes were significantly associated with decreased BC risk in Caucasians2244TNBC is frequently observed in young patients and in patients with larger and higher-grade tumours4851The major strengths of this study were the comprehensive analysis of theTNF-α-308 polymorphism in relation to susceptibility for various BC subtypes and its influence on the clinical features of BC, which will greatly help improve our understanding of the role of TNF-α in BC pathogenesis. The modest sample size of each subtype, which caused suboptimal statistical power, is the main limitation of this study; however, this could not be avoided. In conclusion, the present study shows that the TNF-α-308G > A polymorphism is not associated with BC risk but is associated with distant tumour metastasis in TNBC. This association might be mediated by the constitutively higher expression of tmTNF-α and/or sTNF-α in patients with the TNF-α-308A allele, promoting tumour growth through metastasis. Our results also confirm that targeting TNF-α suppresses TNBC progression.−, PR− and Her2−), Her2+ , luminal A, and luminal B . The controls were randomly selected based on a physical examination in the same region during the same period as patient recruitment. The selection criteria included no history of cancer and frequency matching to cases by age. At recruitment, demographic information and clinical characteristics of each participant were collected. Informed consent was obtained from all participants. This study was conducted in accordance with the approved guidelines of the Institutional Review Board of the Cancer Hospital, Shandong Academy of Medical Sciences and Beijing Chao-Yang Hospital, Capital Medical University.This case-control study included 768 patients with constitutive BC and 565 cancer-free controls. All subjects were unrelated ethnic Han Chinese women. Patients were recruited from January 2010 to December 2013 at the Cancer Hospital, Shandong Academy of Medical Sciences and Beijing Chao-Yang Hospital, Capital Medical University and had been diagnosed with histologically confirmed BC. In this study, we classified the BC subtypes as TNBC  site were analysed using a TaqMan genotyping platform . The PCR primers were 5′-GGC CAC TGA CTG ATT TGT GTG T-3′ and 5′-CAA AAG AAA TGG AGG CAA TAG GTT-3′, and the probe sequences were 5′-FAM-CCC GTC CCC ATG CCC CTC-BHQ-3′ and 5′-TET-CTG AAC CCC GTC CTC ATG CCC-BHQ-3′. For quality control, genotyping was performed by researchers blinded to the case or control status, and a 10% random sample of cases and controls was genotyped twice by different persons; the reproducibility was 99.0%.We used three electronic databases  to identify relevant publications through December 2014 using key words related to the TNF-α gene polymorphism in combination with BC. The literature search was limited to English language publications on human studies. The reference lists of the relevant articles, reviews and editorials were also screened to find all additional eligible studies.Studies were included in this meta-analysis if they met the following criteria: (1) case-control study design comparing TNF-α-308 polymorphism in BC cases and cancer-free controls; (2) cases were diagnosed histologically; (3)genomic DNA was isolated from peripheral blood leukocytes; (4) the study estimated the association between the TNF-α-308 polymorphism and BC risk; (5) sufficient data existed for calculating odds ratios (ORs) and corresponding 95% confidence intervals (CIs); and (6) the allele frequencies in the control group met HWE. If the same subject group was published in more than one study, the most complete study was selected for this pooled meta-analysis.All information extracted from the included studies, including the first author, the year of publication, the ethnicity of the subjects, the number of subjects and genotype frequencies in the case and control groups, and the genotyping method, was double-checked and independently extracted from each publication by two investigators. Disagreements were resolved through discussion.2 tests were used to examine the deviation in genotype frequencies from HWE in controls and the differences in demographic variables and genotype distributions for different clinical features of BC. A dominant genetic model was used to estimate the associations between the TNF-α-308 polymorphism and the risk of BC by ORs and 95% CIs, which were calculated by unconditional logistic regression adjusted for age. For the meta-analysis, subgroups of studies based on ethnicity and experimental methods were determined as described in our previous studyP value less than 0.05 was considered statistically significant.χHow to cite this article: Li, H.-H. et al. Tumour Necrosis Factor-α Gene Polymorphism Is Associated with Metastasis in Patients with Triple Negative Breast Cancer. Sci. Rep.5, 10244; doi: 10.1038/srep10244 (2015).


To estimate the relationship between the risk of preeclampsia and two thrombophilia gene single-nucleotide polymorphisms (SNPs), the factor V G1691A SNP and the prothrombin G20210A SNP.A systematic search of the English-language literature up to November 2012 was performed using Medline and EMBASE. Search terms included “preeclampsia,” “thrombophilia,” “factor V Leiden,” “prothrombin gene 20210,” and their combinations.Thirty-seven studies with 5048 preeclampsia patients and 6796 controls were included in the meta-analysis. We found that the prothrombin G20210A polymorphism was associated with an increased risk of all preeclampsia (pooled odds ratio (OR) = 1.81, 95% confidence interval (CI) 1.25–2.63) and severe preeclampsia . Meanwhile, the pooled OR for the association between factor V Leiden and all preeclampsia was 1.60 (95%CI 1.28–2.00) and 2.45 (95%CI 1.63–3.69) for the cases of severe preeclampsia.This meta-analysis supports that the factor V G1691A SNP and the prothrombin G20210A SNP are associated with an increased risk for both preeclampsia overall and severe preeclampsia. Preeclampsia (PE), a complex disorder of pregnancy, is a major cause of maternal mortality worldwide, and is responsible for 15% to 20% of maternal mortality in developed countries Despite many studies conducted previously, the etiology of PE remains unclear. Many hypotheses have been presented to explain the development of PE, including maternal and fatal genetic and environmental factors, classified into four categories: 1) immune maladaptation, 2) placental ischemia, 3) oxidative stress, and 4) genetic susceptibility These inconsistent findings may be due to the differences in sample size, study design, ethnicity, and other aspects A computerized search of the English-language literature up to November 2012 was performed using Medline and EMBASE. We used the following search terms and the combinations thereof: “preeclampsia,” “thrombophilia,” “factor V Leiden,” and “prothrombin gene 20210.” The inclusion and exclusion criteria were as follows: (a) published studies with full text about the relationship between FVL or prothrombin G20210A and preeclampsia; (b) case-control studies; (c) studies with sufficient data to calculate odds ratio (OR) and 95% confidence interval (CI); (d) Reviews, letters, and single case reports were excluded; (e) if one group of patients was used in two or more studies, the study with the largest number of patients was selected.Two investigators performed the literature search and screened the studies independently according to the inclusion and exclusion criteria. Agreements were reached after discussions if discrepancies occurred.When all the eligible studies were included, we extracted the following information for each study: first author’s name, year of publication, study design, country, ethnicity , and number of cases and controls for each FVL or prothrombin gene polymorphism genotype. In terms of disease definition, the criteria of the American College of Obstetricians and Gynecologists In several studies, the authors classified the patients into two categories according to the severity of disease: severe PE group and not severe PE group while only severe preeclampsia patients were recruited into case group in some other studies. However, no similar classification was made in the remaining studies, in which each individual diagnosed as preeclampsia (not necessarily as severe preeclampsia) was included. To explore whether genetic thrombophilias increase the risk for preeclampsia in general or only for severe preeclampsia, we conducted two separate analyses: one involving all patients with preeclampsia (severe and not severe), and another separate analysis of severe preeclampsia in those studies that identified these cases separately.Q-statistic test and the I-squared statistical test. A fixed-effect model using the Mantel-Haenszel method was applied to calculate the pooled OR and 95% confident intervals (95%CI) if no or small heterogeneity was found. Otherwise a random-effect model using the DerSimonian-Laird method was used. Meta-regression was conducted to explore the source of heterogeneity, and influence analysis was performed when necessary. Studies with any zero cells were adjusted by adding 0.5 to each cell of the 2×2 table. Publication bias was assessed by funnel plot and Egger’s test. All analyses were performed using Stata 11.0 software .Heterogeneity among studies was measured by the Based on our search criteria, a total of 42 studies were included in this analysis. Two studies Q statistic for heterogeneity of 8.10 (p = 0.920). No heterogeneity was found (I2 = 0.0%), and the fixed model was used. These results are shown graphically in p heterogeneity of 0.390. The results are shown in The effect of the PT 20210 polymorphism and severe preeclampsia is shown in p = 0.251, I2 = 15.5%) when a fixed model was applied  was found between studies. Thus, a random model was used for which the pooled OR was 2.45 (95% CI 1.63–3.69). The results are shown graphically in Seventeen studies with 1814 patients and 2721 controls evaluated the association between FVL and severe preeclampsia, as shown in Preeclampsia is multifactorial gestational vascular complication that greatly threatens the life of pregnant women and their fetuses. The etiology of preeclampsia remains complex. Early studies reported that thrombophilia genes were associated with a hypercoagulable state In our meta-analysis, a 2-fold increased risk of preeclampsia and an approximately 3-fold increase for severe preeclampsia are observed when the prothrombin 20210A allele exists. After the exclusion of one study These results are inconsistent with those of Lin and August Q-statistic test suggested a mild heterogeneity between studies. The funnel plot was asymmetric, and the Egger test suggested obvious evidence of small-study bias. These results are similar to those of former meta-analyses Our meta-analysis also indicates that FVL is associated with an increased risk of preeclampsia. The combined OR for FVL associated with all preeclampsia was 1.60 (95%CI 1.28–2.00), and the I2 = 56.1%, p = 0.003). This unexplained heterogeneity may be due to the wide variation of the prevalence of the FVL in the control populations. In most studies, the prevalence of FVL in the control group ranged from 2% to 5%, including all of the larger studies in this meta-analysis, All of them reported ORs much greater than 1. On the contrary, seven studies I2 = 25.1%, p = 0.212 versus I2 = 56.1%, p = 0.003). The funnel plot showed no significant evidence of publication bias.When the studies assessing the relationship between FVL and severe preeclampsia were involved in meta-analysis, there was a 2.45-fold increase in risk. High heterogeneity was found among all studies Click here for additional data file.


We read with great interest the recently published article by Seo et al.  in whichFirst, NLR, which integrates the detrimental effects of neutrophilia (an indicator of inflammation) and lymphopenia (an indicator of physiologic stress), has emerged as a useful prognostic marker in many studies, as well as studies evaluating patients with idiopathic sudden sensorineural hearing loss , 3. In tSecond, although presence of diabetes mellitus (DM) was also specified as another exclusion criterion, as understood from the glucose values presented in Tables 2 and 3 of the original study, some patients with impaired fasting glucose (IFG) and/or impaired glucose tolerance (IGT) were also included in the original study. As is known, NLR values can easily be affected by the existence of inflammation in patients with DM and prediabetes (IFG/IGT) . In a stThird, in the original study, it is noted that binary logistic regression model was used to evaluate these values as independent risk factors for recovery of ISSNHL . In this model authors evaluated lymphocyte, monocyte, NLR, PLR, and glucose. However, before performing the binary logistic regression analysis, univariate logistic regression analysis and correlation analysis should be performed due to the probable associations among the lymphocyte, NLR, and PLR; then significant parameters derived therefrom should be evaluated by binary logistic regression analysis. As is known, correlation of parameters with each other is very important in binary logistic regression analysis.In conclusion, we think that the findings of the present study may be affected by confounding factors mentioned above and additional statistical analysis taking all these factors into account should be performed. In such studies aiming at determining a disease-specific cut-off value, being more discerning in patient selection will provide more accurate results. In this way, further information can be obtained about the association between NLR and sudden sensorineural hearing loss.