Associations between interleukin 6 (IL-6) polymorphisms and Alzheimer’s disease (AD) remain controversial and ambiguous. The aim of this meta-analysis is to explore more precise estimations for the relationship between IL-6-174 G/C and -572 C/G polymorphisms and risk for AD. Electronic searches for all publications in databases PubMed and EMBASE were conducted on the associations between IL-6 polymorphisms and risk for AD until January 2012. Odds ratio (OR) and 95% confidence intervals (CIs) were calculated using fixed and random effects models. Twenty-seven studies were included with a total of 19,135 individuals, involving 6,632 AD patients and 12,503 controls. For IL-6-174 G/C polymorphism, the combined results showed significant differences in recessive model . As regards IL-6-572 C/G polymorphism, significant associations were shown in dominant model  and in additive model . In conclusion, genotype CC of IL-6-174 G/C and genotype GG plus GC of IL-6-572 C/G could decrease the risk of AD. As a major cause of cognitive decline in the elderly, Alzheimer’s disease (AD) currently affects 20–30 million individuals worldwide. It is estimated that the number of people likely to be affected will be triple over the next 50 years throughout the world. Two independent variants in the promoter region of IL-6 promoter (−174 G/C and −572 C/G) have been detected. Previous studies concerning association between IL-6-174 G/C polymorphism and risk of AD are limited and rather conflicting. The second IL-6 polymorphic exchange −572 C/G  is not commonly analyzed as −174 G/C. However, previous findings suggested that −572 C/G polymorphism might affect the transcription rate of the IL-6 gene, and furthermore influence plasma levels of acute proteins such as fibrinogen and C-reactive protein. Two reviewers independently searched studies on the associations between IL-6 polymorphisms and AD. Published studies were identified through a computerized search of PubMed, MEDLINE, EMBASE, and the Cochrane Library in any language up to January 2012. The keywords were as follows: Alzheimer’s disease, interleukin and polymorphism or variant or genotype or SNP. The references of all identified publications were searched for additional studies, and the PubMed option “Related Articles” was also used to search for potentially relevant papers. We only included published articles written in English.Two reviewers independently identified potential relevant studies and evaluated each trial based on predetermined eligibility criteria. Studies were included if they met the following criteria: (1) the study reported original data from case-control studies; (2) the outcome had to be AD; (3) at least two comparison groups (AD patient group vs. control group) and the number of subjects possessing genotype in the AD and control groups were available; and (4) in the case of multiple publications from the same study group, the most complete and recent results were used.After excluding the overlap studies and including the additional ones, this meta-analysis covered a total of 22 articles on IL-6-174 G/C polymorphism and 5 articles on IL-6-572 C/G polymorphism. Two authors extracted the data independently and in duplicate. The following data were extracted from the eligible studies: the first author’s last name, year of publication, country of origin, ethnicity, and numbers of genotyped cases and controls. Any disagreement was adjudicated with a third author.2I statistic interpreted as the proportion of total variation contributed by between-study variation. If there was a statistical difference in terms of heterogeneity , a random-effect model was selected to combine the data. Otherwise, a fixed-effect model was employed. The significance of the pooled OR was determined by the Z-test, and two-tailed P<0.05 was considered as statistically significant. Relative influence of each study on pooled estimates was assessed by omitting one study at a time for sensitivity analysis. Evidence of publication bias was determined by visual inspection of the funnel plot. All statistical analyses were performed with Revman 5.0 .The strength of the association between the IL-6 polymorphism and AD was measured by odds ratio (ORs) with 95% confidence intervals (CIs). We explored the allele comparison, as well as dominant model and recessive model. Heterogeneity among studies was examined with In this article, the associations of IL-6 polymorphisms with AD susceptibility were investigated using meta-analysis in a wide range of populations Figure 1P = 0.53). There was a significant difference between-study heterogeneity (I2 = 67%). No significant associations were also found for IL-6-572 C/G allele . Heterogeneity was detected (I2 = 68%).The association between the IL-6 polymorphisms and AD were showed in  and P = 0.0003, 2 = 65%). No associations were found in dominant model or additive model. For IL-6-572 C/G polymorphism, significant decreased risk was found in additive model , indicating that the results of this meta-analysis were stable.The main finding of this meta-analysis is that genotype CC of IL-6-174 G/C and genotype GG plus CG of IL-6-572 C/G are potential protective factors for developing and progressing AD.The IL-6 gene polymorphisms are widely investigated in relation to the risk of AD. However, the results from these studies were ambiguous, for their small sample size and unified ethnicity. Although two meta-analysis studies have focused on −174 G/C polymorphism recently, none of them could give us results with all publications. Because heterogeneity was found among the studies, we employed random-effect model. Then, a sensitivity analysis was performed by removing one study for each time and re-running the model to determine the effect on the overall estimate. The estimates changed quite little, strengthening the results from this meta-analysis. No publication bias was shown, also strengthening our results.As inflammation is involved in the pathogenesis of AD, IL-6 has been implicated in both the development and progression through accumulation of acute phase proteins in plaques and elevation of amyloid precursor protein synthesis in experimental and human AD. This meta-analysis has pooled all the available results from the case–control studies, which has significantly increased the statistical power. However, some limitations of this meta-analysis should be acknowledged. First, AD is a complex disease that results of combined effects of multifactor, including inherited and environmental factors. Some environmental factors may strongly influence the development of AD. Lacking of considering these factors may affect the significance for the independent role of IL-6 polymorphisms in AD development. Second, the network of cytokines is complex and variants of other pro-inflammatory cytokines may exert their complex and interacting functions with each other. No regard of these factors may cause serious confounding bias. Third, the number of current studies on IL-6-572 C/G is relative small. Thus, investigations involving more subjects of different races are needed to confirm the effect and then another meta-analysis should be necessary for a more reliable evaluation on their associations. Fourth, the heterogeneity observed in the relationship between −174 G/C polymorphism and AD, even when analyzed by ethnicity and a sensitivity analysis. Finally, we reviewed only published studies.In conclusion, this meta-analysis detected significant associations between IL-6-174 G/C and −572 C/G polymorphisms and AD. Genotype CC of IL-6-174 G/C and genotype GG plus CG of −572 C/G could decrease the risk of AD. To better understand the potential mechanism for AD in humans, large well-designed cohort studies are needed to confirm these associations and further researches should be carried out to explore the effect of genetic networks, environmental factors, individual biological characteristics and their mutual interactions.


Therefore, we examined whether the 5-HTTLPR gene and 5-HT1B gene G861C polymorphism are susceptibility factors for a specific subtype of alcoholism, antisocial alcoholism in Han Chinese in Taiwan.Several studies have hypothesized that genes regulating the components of the serotonin system, including n = 120) and antisocial non-alcoholism (AS-N-ALC) group (n = 153)] and 191 healthy male controls from the community. Genotyping was done using PCR-RFLP.We recruited 273 Han Chinese male inmates with antisocial personality disorder (ASPD) [antisocial alcoholism (AS-ALC) group (5-HT1B G861C polymorphism between the 3 groups. Although AS-ALC group members more frequently carried the 5-HTTLPR S/S, S/LG, and LG/LG genotypes than controls, the difference became non-significant after controlling for the covarying effects of age. However, the 5-HTTLPR S/S, S/LG, and LG/LG genotypes may have interacted with the 5-HT1B G861C C/C polymorphism and increased the risk of becoming antisocial alcoholism.There were no significant differences in the genotypic frequency of the 5-HTTLPR gene nor the 5-HT1B G861C polymorphism alone is a risk factor for antisocial alcoholism in Taiwan’s Han Chinese population, but that the interaction between both genes may increase susceptibility to antisocial alcoholism.Our study suggests that neither the  Alcoholism  is a complex, heterogeneous disorder that is influenced by multiple genes as well as sociocultural factors ,2. Twin ADH1B*2 allele against alcoholism vanished in antisocial alcoholics [ALDH2*2 allele might disappear in individuals with antisocial personality disorder (ASPD) and carrying the MAOA-uVNTR 4-repeat allele [DRD2 TaqI A1 (A1A1 or A1/A2) and 5-HTTLPR S/S genotype [serotonin (5-HT) receptor 1B A-161T from other subtypes of alcoholics [Cloninger 1987) identified two subtypes of alcoholism. Type I alcoholism is characterized by late onset, more psychological dependence, and anxious/depressed personality traits. Individuals with type II alcoholism are characterized by early onset, a higher familial risk for alcoholism, and antisocial personality traits 987 ident. We showcoholics , and the5-HTT-linked promoter region (5-HTTLPR), which contains an insertion-deletion of two 22-bp repeats in a region of the promoter with multiple 22- to 23-bp repeats [5-HTTLPR displayed reduced transcriptional activity [5-HTTLPR polymorphisms were also correlated to the function of serotonin transporter both in vivo imaging study [L/L) had an increase in the availability of raphe serotonin transporter [S/L and S/S genotypes were also lower than in subjects with the L/L genotype in postmortem brain study [LA) has been reported to have higher activity than the long variant with a guanine at rs25531 (LG) [XL variants with 15, 18,19, 20 or 22 repeats with unknown function [5HTTLPR into S and L allele may not be sufficient and the association between the tri-allelic 5-HTTLPR polymorphism (SLALG) and the risk of alcoholism or antisocial alcoholism remains unknown.Numerous studies have reported reduced brain serotonin  function in patients who are similar to antisocial alcoholics. Lower cerebrospinal fluid concentration of the 5-HT metabolite 5-HIAA (5-hydroxyindoleacetic acid) has been found in early-onset antisocial alcoholics , impulsi repeats , affects repeats . In vitractivity  as well activity . 5-HTTLPng study  and in png study . Human Snsporter . Serotonin study . Althougin study -26, thisin study ,27,28. Iin study . The lon531 (LG) ,31. Furt531 (LG) ,32-35. Tfunction ,32-35. H5-HT1B gene has been associated with alcohol intake and aggression in animal and human studies. An alcohol preference locus has been mapped in mice to the region where 5-HT1B is located [5-HT1B receptor knockout mice were found to have increased aggressive behavior [5-HT1B, the results of association studies that used the G861C polymorphism remain inconsistent [5-HT1B G861C with alcoholism [5-HT1B 861C allele also had 20% fewer 5-HT1B binding sites in the prefrontal cortex, which was the same as the alcoholism group [The serotonin 1B receptors (5-HT1B) are located on presynaptic and postsynaptic terminals and function as both autoreceptors and heteroreceptors that mediate the release of serotonin . The 5-H located . 5-HT1B behavior  and incrbehavior . In humansistent . Some stcoholism  or antiscoholism . People sm group ,44. Howesm group -47. The The prevalence rate of ASPD and antisocial behaviors in individuals with alcohol use disorders ranges from 20% to 33% -50. Beca5-HTTLPR polymorphism and 5-HT1B G861C polymorphism are associated with AS-ALC in the Han Chinese population in Taiwan, and (2) assess the serotonergic genetic interaction between 5-HTTLPR and 5-HT1B G861C polymorphisms and antisocial alcoholism.The aims of current study is to (1) investigate whether the tri-allelic n = 173) and a prison in Tainan (southern Taiwan) (n = 100). The research protocol was reviewed and approved by the Taiwan Ministry of Justice, which permitted us to collect blood samples from the participants. The procedures were fully explained to each participant before they gave written informed consent; however, there were no advocates for the inmates present when they were asked to sign the consent form. Participants were told that they were free to drop out of or to join the study at any time. We offered several compensatory benefits: 1) a free physical examination, 2) liver and renal function tests, and 3) and free treatment or assistance in transferring those with psychiatric or physical illnesses to hospitals. Each participant was initially evaluated by an attending psychiatrist and then interviewed by a psychologist well-trained in using the DSM-IV and the Modified Chinese Version of the Modified Schedule of Affective Disorder and Schizophrenia-Life Time (SADS-L) [n = 120) or the AS-N-ALC group (n = 153), respectively.This study was approved by the Institutional Review Boards for the Protection of Human Subjects at the Tri-Service General Hospital and National Cheng Kung University Hospital. Because of the low prevalence (0.1%) of ASPD in Han Chinese in Taiwan , it is d(SADS-L) ,54. The (SADS-L) ,55-57. PThe healthy control group included 191 male volunteers recruited from the community. The Chinese version of the SADS-L was used to screen their psychiatric conditions. All controls were free of present and past major and minor mental illness , and none had a family history of psychiatric disorders among their first-degree relatives.5-HTTLPR functional polymorphisms were genotyped using polymerase chain reaction plus restriction fragment length polymorphism (PCR-RFLP) analysis [L allele were determined using a modified protocol [G861C polymorphism of the 5-HT1B gene was genotyped using a modified protocol [Twenty milliliters of blood was drawn from each participant, and standard methods were used to extract genomic DNA from the leukocytes. The genotypes of analysis . The A/Gprotocol . The G86protocol .5-HTTLPR and 5-HT1B G861C polymorphisms between three groups was calculated using the Pearson χ2 test (2-tailed). The 5-HTTLPR polymorphisms in AS-ALC and AS-N-ALC groups were further stratified by the 5-HT1B G861C polymorphism using the χ2 tests. Logistic regression analysis was used to examine the main effect and gene-to-gene interaction of the 5-HTTLPR and 5-HT1B G861C polymorphisms for the risk of antisocial alcoholism. Both the SA and the SG alleles were recorded as the S allele because there is no functional difference in the SNPs [S and LG carrier had lower function and expression of serotonin transporter [5-HTTLPR polymorphisms into 2 groups, low-functional (S/SS/LGLG/LG) and high-functional (S/LALG/LALA/LAS/XLLA/XLLG/XL), for data analysis. There was only 8 participants carried the LALA genotype in our study, so we did not separate the homozygous LA carries from heterozygous LA carriers into another group. SPSS for Windows 17.0  was used for all statistical analyses.One-way analysis of variance (ANOVA) and the Bonferroni Post Hoc Test were used to determine the mean age differences between the subgroups. The difference in genotype frequency of the SNPs . Becausensporter -20,60, wn = 464) had a power of 0.58 to detect a small effect (effect size = 0.1), of 0.99 to detect a medium effect (effect size = 0.3), and of 1.00 to detect a large effect (effect size = 0.5) of genotype distributions.The power estimation was calculated using G-power 3.1 software ,62. Our P = 0.004) and AS-N-ALC groups . The distribution of the 5-HTTLPR and 5-HT1B G861C genotypes was not significantly different between the three groups  in the AS-ALC group than in the control group  and the 5-HT1B G861C C/C genotype may increase the risk of becoming an antisocial alcoholic  = 6.22 and 5.76, respectively)  controls. Because ASPD has been linked to the 5-HTTLPR polymorphism [5-HTTLPR polymorphism instead of a bi-allelic polymorphism. All these factors may have caused results different from those of other reports.The interaction between the l cortex ,44 and tal group ,31, our morphism -65, we tmorphism ,26, the 5-HT1B G861C polymorphism. A previous study [5HT1B 861 C allele varied widely among different ethnic groups, ranging from 20% in African-Americans to 36% in Hispanics and 42% in Asians. In our current study, the frequency of the 5-HT1B C allele  was higher than in previous reports. We also added an AS-N-ALC group to decrease the effect of ASPD. Different populations and ethnicities may also partially explain our different results.We found no association between antisocial alcoholism and the us study  showed t5-HTTLPR and 5-HT1B genes in the three groups. It seems that the antisocial non-alcoholics and healthy controls had a similar genotype distribution of these serotonin-related genes, one different from that of the antisocial alcoholics. We may speculate that other serotonin-related genes have the same distribution between individuals with ASPD and the general population. In addition, serotonin may modulate the reinforcement processes, possibly through complex interactions with the mesolimbic dopamine system [dopamine D2 receptor gene interacted with the 5-HTTLPR gene in antisocial alcoholics with specific personality traits. Hence, additional studies to explore the associations among the 5-HTTLPR, the 5-HT1B gene, 5-HT metabolic enzyme-related genes, and dopamine-related genes will be required to study the pathogenesis of AS-ALC.Our past study  showed te system ,67 and te system -70. The e system ,70,71. Fe system , and thee system . We havee system  that the5-HTTLPRLA/LA genotype, and about 3.8% with the 5-HTTLPR XL allele, which is different from other ethnic groups [5-HTTLPRS/LALG/LALA/LAS/XLLA/XLLG/XL genotypes into a high-functional group and those with the S/SS/LGLG/LG genotypes into a low-functional group for statistical analysis according to past study results [5-HTTLPR XL allele remains unknown, one study had speculated that it may function as an L allele [42.Our study has several limitations. First, the study populations in the AS-ALC and AS-N-ALC groups were small. In addition, the cell size with the significant finding was also small, which may cause the risk of false positive results. Larger study populations are needed. Second, there were few participants with the c groups . It may  results -20,60. AL allele . HoweverL allele . FinallyL allele . It is n5-HTTLPR and 5-HT1B G861 polymorphisms. Additional studies to examine the 5-HTTLPR and 5HT1B G861C gene polymorphisms and their interaction effects with the dopaminergic gene in antisocial alcoholism are required.Our results supported the notion that antisocial alcoholism is associated with the interaction between the Obtained.This study was approved by the Institutional Review Boards for the Protection of Human Subjects at the Tri-Service General Hospital and National Cheng Kung University Hospital.5-HTTLPR, Serotonin transporter; 5-HT1B, Serotonin 1 B receptor; ASPD, Antisocial personality disorder; AS-ALC, Antisocial alcoholism; AS-N-ALC, Antisocial non-alcoholism.The authors declare no conflict of interests.The author, TYW wrote the first draft of this manuscript and designed this study with the author RBL. The authors, SYL and Y-HC managed the statistical analyses. The authors, SLC, SHC, CHC and CLC managed the lab work. The authors, TYW, SYH, NST, IHL, TLY, YKY and RBL managed the patients recruitment. All authors read and approved the final manuscript.


Association studies of serotonin transporter gene SLC6A4 I/S polymorphism and irritable bowel syndrome (IBS) have shown inconsistent and contradictory results among different populations. In the present study, meta-analysis was performed to evaluate the association between SLC6A4 I/S polymorphism and IBS susceptibility.Systemic assessment was performed for the published studies based on the association of SLC6A4 I/S polymorphism and IBS risk from PubMed (Medline), EMBASE search. A meta-analysis was done to appraise the said association. Pooled odds ratios (ORs) and 95% confidence intervals (CIs) were calculated for allele contrast, homozygous, heterozygous, dominant and recessive genetic model.A total of twelve studies comprising 2068 IBS cases and 2076 controls were included in this meta-analysis. Overall, no significant results were obtained for S allele carrier  Co-dominant , . Similarly, dominant  and recessive  genetic models did not demonstrate risk. In the subgroup population based analysis, reduced risks were found in American  and Asian  population. However, no risk was observed in European population.This investigation clearly demonstrates that SLC6A4 (Ins/Del) polymorphism is associated with reduced risk of IBS in American and Asian population. However, future well-designed studies with stratified case control and biological characterization will be needed to validate this finding. Irritable bowel syndrome (IBS) is a predominant and common chronic gastrointestinal (GI) disorder, associated clinical manifestations are recurrent abdominal pain or discomfort, swelling, and altered bowel function, such as constipation, irregular stool frequency and alternation between these two symptoms ,2. The pSeveral polymorphisms have been discovered in this gene, among them polymorphism of a 44-base pair insertion (Ins)/deletion (Del) in the 5’-ﬂanking promoter region (5-HTTLPR), which results in either a long (l) or short (s) transcript has been identified as important one. Homozygous (S/S) and heterozygous (I/S) genotypes have been shown to be associated with lower transcriptional activity compare with of the homozygous long (I/I) genotyping in transfected cell line, and thus lower levels of SERT mRNA and consequently 5-HT reuptake ,11. The Having known the functional significance of the serotonin signaling in the gastrointestinal tract, it has been considered that SLC6A4 polymorphism may be a potential susceptibility factor for IBS. Till now, a number of studies have been conducted in different populations but the results are conflicting rather than conclusive, because some studies report association between IBS and SLC6A4 (Ins/Del) polymorphism, whereas others did not show statistical relationship between them -24. IncoThis meta-analysis was conducted according to the standard guidelines of observational studies in epidemiology .We performed a PubMed (Medline), EMBASE search covering all research papers published with a combination of the following key words: “SERT, SLC6A4, 5-HTT gene, Serotonin transporter gene (polymorphism OR mutation OR variant) AND irritable bowel syndrome or IBS (last updated on March 2013). The references provided in the retrieved publications were also reviewed in case any relevant study was missed. Potentially relevant genetic association studies were evaluated by examining their titles and abstracts, and all published studies matching with the above said key words were retrieved.In order to minimize heterogeneity and facilitate the proper interpretation of our results, studies included in the current meta-analysis had to meet the following criteria: a) evaluated the association between Ins/Del I>S and IBS risk, b) used a case-control design, c) recruited pathologically confirmed IBS patients and healthy controls, d) have available all three genotype frequencies in cases and controls, e) published in English language. The major reasons for exclusion of studies were, a) overlapping data, b) case-only studies and review articles, and c) genotype frequencies or number not reported. The supporting PRISMA 2009 Flow Diagram (showing identification and selection of studies) is available as supporting information; see For each publication, the methodological quality assessment and data extraction were independently abstracted in duplicate by two independent investigators using a standard protocol and data-collection form according to the inclusion criteria listed above to ensure the accuracy of the collected data. In case of disagreement on any item of the data, the problem was fully discussed to reach a consensus. Characteristics abstracted from the studies included the name of the first author, year of publication, the country of origin, the number of cases and controls, types of study, genotype frequencies. These characteristics were important for representing the relevant published studies on the said topic. Also, these characteristics reflect the important features of the included published reports and help in retrieval and utilization of these reports for relevant or similar studies.2 statistics was also used to quantify inter-study variability and larger values indicated an increasing degree of heterogeneity , so random effect model was applied. However, publication bias did not exist . We obseIn European population, random effect model was applied in three  genetic models based on heterogeneity. However, no publication bias was detected . No signWhile accessing the Asian population, random effect models were applied for heterogeneity and no publication bias was detected . The evaDespite recent advancements in IBS diagnosis, the prognosis of patients with IBS remains dismal. It is well known that IBS susceptibility is determined not only by the infectious agents, psychosocial and environmental factors, genetic factors also play an important role in their pathogenesis ,33. In tOverall analysis of pooled results indicated that SLC6A4 (Ins/Del) polymorphism has no significant association with the IBS risk. Whereas, when we stratified the selected studies by the participants’ region or population, we detected a reduced effect of the polymorphism on IBS risk in American and Asian studies. Previously, Van Kerkhoven et al. performed the meta-analysis and reported no association between SLC6A4 (Ins/Del) polymorphism and IBS susceptibility . HoweverAlterations in 5-HT signaling, such as, release or reuptake may contribute to chaotic GI function and hypersensitivity reported in IBS patients . EarlierIn addition, IBS pathogenesis is considered to be multifactorial and single genetic variant is usually insufficient to predict the risk of this disease. One important characteristic of this gene polymorphism is that their incidence may vary substantially between different populations and/or ethnicities. Heterogeneity between the studies is very common in the meta-analysis of genetic association studies. While pooling all the eligible data we found inter-study heterogeneity. But after subgroup analysis by participant’s region or population types, the heterogeneity was effectively decreased. Therefore, it can be assumed that the relatively heterogeneity mainly results from differences of population and/or ethnicity. Also, other factors might be involve in this heterogeneity, such as genotype distributions of SLC6A4 (Ins/Del) locus varied between different population and environmental exposures in different case-control studies were not investigated; possibly, these may also influence the genetic susceptibility. Certain limitations were associated with our study, which may affect the results and should be addressed in future studies. For e.g., First, studies published in English language were included; studies indexed by the selected electronic databases were included for the data analysis; it is possible that some relevant articles published in other language and indexed in other databases may have missed. Second, these results are based on unadjusted ORs.In conclusion, results of this meta-analysis clearly indicate that SLC6A4 (Ins/Del) polymorphism is associated with reduced risk of IBS in American and Asian population. The importance of this polymorphism as a predictor of the risk of IBS is probably limited and the screening utility of this genetic variant in asymptomatic individuals may not be warranted. The heterogeneity of the IBS population poses challenges for investigators focused on its pathogenesis and therapy. Well-designed large scale association studies incorporated with environmental factors in different populations are needed to validate this association as identified in the present meta-analysis and to investigate the potential gene-gene and gene-environment interaction between SLC6A4 (Ins/Del) polymorphism and susceptibility to IBS.Checklist S1PRISMA 2009 Checklist.(DOC)Click here for additional data file.Figure S1PRISMA 2009 Flow Diagram: Showing identification and selection of studies for the meta-analysis.(TIF)Click here for additional data file.


Variants of UCP2 and UCP3 genes have been reported to be associated with obesity, but the available data on the relationship are inconsistent. A meta-analysis was performed to determine whether there are any associations between the UCP2 -866G/A, Ala55Val, and UCP3 -55C/T polymorphisms and obesity susceptibility.The PubMed, Embase, Web of Science and CNKI, CBMdisc databases were searched for all relevant case-control studies. The fixed or random effect pooled measure was determined on the bias of heterogeneity test among studies. Publication bias was examined by the modified Begg's and Egger's test.Twenty-two published articles with thirty-two outcomes were included in the meta-analysis: 12 studies with a total of 7,390 cases and 9,860 controls were analyzed for UCP2 -866G/A polymorphism with obesity, 9 studies with 1,483 cases and 2,067 controls for UCP2 Ala55Val and 8 studies with 2,180 cases and 2,514 controls for UCP3 -55C/T polymorphism. Using an additive model, the UCP2 -866G/A polymorphism showed no significant association with obesity risk in Asians . In contrast, a statistically significant association was observed in subjects of European descent . But neither the UCP2 Ala55Val nor the UCP3 -55C/T polymorphism showed any significant association with obesity risk in either subjects of Asian  or of European descent .Our meta-analysis revealed that the UCP2 -866G/A polymorphism may be a risk factor for susceptibility to obesity in subjects of European descent, but not in individuals of Asian descent. And our results did not support the association between UCP2 Ala55Val, UCP3 -55C/T polymorphisms and obesity in the populations investigated. This conclusion warrants confirmation by further studies. Uncoupling proteins, which comprise five UCP homologues (UCP1-UCP5), belong to the family of mitochondrial transporter proteins that uncouple oxidative phosphorylation from ATP synthesis and release excess energy as heat However, the impact of UCP2 and UCP3 polymorphisms on obesity susceptibility is still under debate. Contradictory results have been reported. To further examine their potential role in influencing obesity susceptibility, we performed a meta-analysis on eligible case-control studies to estimate their effects in populations of Asian and European descents.We performed an exhaustive search on studies that examined the association of the UCP2 and UCP3 gene polymorphisms with obesity. Data were collected from the PubMed, Embase, Web of Science and CNKI, CBMdisc databases and completed on September, 2012. We searched the articles using the search terms UCP2; UCP3; uncoupling protein 2; uncoupling protein 3; variant; polymorphism in combination with obesity. Further, we reviewed all abstracts obtained from our search for relevance; manually reviewed bibliographies and review articles for additional citations and obtained the full text of all potentially relevant articles. All searches were conducted independently by two investigators.(1) A case-control study; (2) numbers in case and control groups reported each allele or genotype; (3) sufficient published data to calculate an odds ratio (OR) with 95% confidence interval (CI).(1) review articles; (2) case reports; (3) abstracts and editorials; (4) reports with incomplete data; (5) studies based on pedigree data; (6) genotype distribution of the controls deviated from Hardy – Weinberg equilibrium (HWE).Two investigators independently extracted data, discussed disagreements, and reached consensus on all items. The following information was extracted from each study: the first author's name, year of publication, ethnic origin of the studied population, study design, genotyping methods, number of cases and controls, available allele and genotype frequencies information, and OR with 95% CI. Not all the papers reported the necessary statistics directly so we had to transform and estimate odds ratio from the reported data as necessary, and we did not define any minimum number of patients for a study to be included in our meta-analysis.P<0.05. The strength of association between the UCP2 -866G/A, Ala55Val and UCP3 -55C/T polymorphisms and obesity was assessed by calculating OR with 95% CI in the additive, dominant, recessive and co-dominant models, respectively. The significance of the pooled OR was determined by the Z-test, and P<0.05 was considered statistically significant, stratified analysis was also performed analysis on ethnicity, study design and genotyping methods, respectively. The heterogeneity between the studies was evaluated with x2-based Q statistic and I2 metric. Heterogeneity was considered significant at P<0.05 for the Q statistic and I2 >50% for the I2 metric. The pooled OR was calculated by a fixed effect model (using the Mantel-Haenszel method) or a random effect model (using the DerSimonian-Laird method) according to the heterogeneity among studies et al. P<0.05 was considered representative of statistically significant publication bias). All statistical analyses were conducted by using STATA version 11.0 .HWE of the genotype distribution from the controls was tested by a goodness-of-fit χ2 analysis. The distribution was considered deviated from HWE at A total of 22 published articles The association between the UCP2 -866G/A polymorphism and obesity was investigated in 12 studies with a total of 7,390 cases and 9,860 controls. No significant association between the UCP2 -866G/A polymorphism and the risk of obesity was found using either additive , dominant , recessive , or co-dominant  models . HoweverFor the UCP2 Ala55Val polymorphism, the C allele was found to be not associated with obesity risk using either additive , dominant , recessive  or co-dominant  model . FurtherOur meta-analysis also showed no significant association between the UCP3 -55C/T polymorphism and the risk of obesity in all genetic models  . As the As shown in Funnel plots and Egger's test were performed to assess the publication bias of the literature, as shown in Obesity is a disorder with a strong genetic component and the impact of hereditary factors is estimated to be between 50% and 85% etc. In addition, the FPRP value for the UCP2 -866G/A polymorphism in Europeans is not robust, which suggest the result warrants confirmation by further studies.Here our results indicate that the UCP2-866G/A polymorphisms are obesity susceptibility loci in populations of European, but not Asian population. Ethnic difference between Europeans and Asians provide us a better understanding that structural chromosome variations may affect genes' expression in different ethnic populations and cause ethnic phenotypic diversity or ethnic specific diseases Heterogeneity is potentially a significant problem when interpreting the results of any meta-analysis of genetic association studies etc. Kring SI et alThe results of the present meta-analysis should also be interpreted within the context of its limitations. First, obesity is a complex status involves complex interactions of genes, environment, and behavior. Many studies Despite these limitations, our results indicate that the UCP2 -866G/A polymorphism may be obesity susceptibility loci in populations of European, but not Asian population. However, the UCP2 Ala55Val and the UCP3 −55C/T polymorphisms are not candidate loci for susceptibility to obesity in any ethnic population. Maybe this association is not robust and could be due to chance, and additional larger studies that allow stratification for other gene-gene and gene-environment should also be conducted in future analyses.Figure S1Systematic review flow diagram. We performed an exhaustive search on studies that examined the association of the UCP2 and UCP3 gene polymorphisms with obesity. Data were collected from the PubMed, Embase, Web of Science, CBMdisc and CNKI databases and completed on September, 2012. n, number of studies.(TIF)Click here for additional data file.Figure S2Begg's funnel plot for publication bias test of UCP2 -866 G/A (A), Ala55Val (B) and UCP3 -55 C/T polymorphism (C) and obesity risk.(DOC)Click here for additional data file.Table S1Pooled measures for the association between the UCP2 -866G/A, Ala55Val and UCP3 -55C/T polymorphisms and susceptibility to obesity by study design or methods. HCC, hospital-based case-control study; PCC, population-based case-control study; PCR-RFLP, Polymerase Chain Reaction – Restriction Fragment Length Polymorphism; DASH, dynamic allele-specific hybridization.(DOC)Click here for additional data file.Table S2Egger's publication bias test for the UCP2 -866G/A, Ala55Val and UCP3 -55C/T polymorphisms in obesity risk. It was shown that there was no publication bias for any of the polymorphisms examined .(DOC)Click here for additional data file.


APE1 have been implicated in susceptibility to cancer; however, results from the published studies remained inconclusive. The objective of this study was to conduct a meta-analysis investigating the association between polymorphisms in APE1 and the risk for cancer.APE1 (apurinic/apyrimidinic endonuclease 1) is an important DNA repair protein in the base excision repair pathway. Polymorphisms in APE1 polymorphisms and cancer risk. Odds ratios (ORs) and 95% confidence intervals (CIs) were used to assess the strength of the associations.The PubMed and Embase databases were searched for case-control studies published up to June, 2011 that investigated Two polymorphisms  in 37 case-control studies including 15, 544 cancer cases and 21, 109 controls were analyzed. Overall, variant genotypes (GG and TG/GG) of −656 T > G polymorphism were associated with significantly decreased cancer risk in homozygote comparison , dominant model comparison  and recessive model comparison , whereas the 1349 T > G polymorphism had no effects on overall cancer risk. In the stratified analyses for −656 T > G polymorphism, there was a significantly decreased risk of lung cancer and among Asian populations.APE1 −656 T > G polymorphism has a possible protective effect on cancer risk particularly among Asian populations whereas 1349 T > G polymorphism does not contribute to the development of cancer.Although some modest bias could not be eliminated, the meta-analysis suggests that  Cancer is a multifactor disease caused by complex interactions between environmental and genetic factors . The incAmong DNA repair systems, base excision repair (BER) pathway is responsible for repairing small lesions such as oxidative damage, alkylation, or methylation . AP EndoAPE1 is located on chromosome 14q11.2-q12 and consists of five Exons spanning 2.21 kb [APE1 may confer individuals' susceptibility to cancer. A total of 18 single nucleotide polymorphisms (SNPs) in APE1 have been identified [APE1 and contributed to lung cancer susceptibility [The human tibility . Althougtibility -17. As ttibility . RecentlAPE1 -656 T > G and 1349 T > G polymorphisms on cancer risk; we conducted a meta-analysis of all eligible case-control studies that have been published. Before analysis, we tested whether these two SNPs were in linkage disequilibrium with each other in Europeans and Asians, and the data from HapMap database suggested the two SNPs were not correlated .To reveal a small effect of the polymorphisms on cancer risk, a single study might be underpowered, particularly for studies with relatively small sample size. Meta-analysis is a statistical technique for combining results from different studies to produce a single estimate of the major effect with enhanced precision. It is considered a powerful tool to summarize inconclusive results from different studies. To clarify the effect of the APE1 and cancer risk were retrieved, and their bibliographies were checked for other relevant publications. Studies that were included in the meta-analysis had to meet all of the following criteria: 1) evaluate the APE1 −656T > G and 1349T > G polymorphisms and cancer risk, 2) use a case-control design, 3) contain available genotypes frequency for estimating an odds ratio (OR) with a 95% confidence intervals (CIs), 4) genotype distributions in control consistent with Hardy-Weinberg equilibrium (HWE). Accordingly, abstracts and reviews studies that that did not report genotype frequency were excluded. When more than one of the same patient population was included in several publications, only the most recent or complete study was used in this meta-analysis.A literature search of the PubMed and EMBASE databases  was conducted using combinations of the following terms: "APE1", "APEX", "HAP1", "REF-1", "polymorphism", and "cancer". The search was limited to English language papers. All studies that evaluated the association between polymorphisms of Information was carefully extracted from all eligible publications independently by two of the authors according to the selection criteria. In case of disagreement, a third reviewer assessed the articles. The following items were collected: author name, year of publication, country of origin, ethnicity, source of the controls, genotyping method, cancer type, total number of cases and controls, and genotype distributions in cases and controls.APE1 polymorphisms and cancer risk were measured by ORs with 95%CIs. We examined the association between the APE1 −656 T > G and 1349 T > G polymorphisms and cancer risk in homozygote comparison (GG vs. TT), heterozygote comparison (TG vs. TT), the dominant genetic model (TG/GG vs. TT), and the recessive genetic model (GG vs. TT/TG). Stratified analyses were also carried out by ethnicity, cancer type and source of controls.The strength of associations between P value is greater than 0.05 of the Q-test, which indicates a lack of heterogeneity among studies, the summary OR estimate of each study was calculated by a fixed effects model (the Mantel-Haenszel method) [P values.Heterogeneity assumption was evaluated with a chi-square-based Q-test. If the  method) ; otherwi method)  was perfA total of 37 eligible studies from 32 articles met the inclusion criteria -17,22-47P < 0.001).The frequency of the G allele varied widely across the 8 studies, ranging from 0.44 to 0.64. The average frequency of the G allele in Asian populations was 0.45, which was lower than that in European populations (0.62). The difference was statistical significant  for the main effects. Therefore, we conducted sensitivity analyses to locate the source of heterogeneity, and the results suggested that two studies [P heterogeneity = 0.123, 0.355 and 0.163, respectively for homozygote comparison, dominant and recessive model). The results were not materially altered after exclusion of these studies, suggesting the results of this meta-analysis were stable.Heterogeneity was observed in homozygote comparison . To adjust for this bias, a trim-and-fill method developed by Duval and Tweedie [As shown in Figure  Tweedie  was condP = 0.514).The frequency of the G allele varied widely across the 39 studies, ranging from 0.20 to 0.52. However, no significant difference in the frequency distribution of the G allele between Asian and European populations were observed .As shown in Table APE1 are associated with the risk of cancer; however, the data have yielded conflicting results. In the present study, to derive a more precise estimation of the relationship, we performed a meta-analysis of 37 published studies including 5,139 cases and 5,201 controls for APE1 −656 T > G polymorphism, and 14,222 cases and 19,746 controls for APE1 1349 T > G polymorphism. Our meta-analysis indicated that variant genotypes (GG and TG/GG) of −656 T > G polymorphism was associated with a significant decrease in the overall risk of cancer, especially for lung cancer and Asians. However, the 1349 T > G polymorphism did not appear to have significant influence on the overall risk of cancer.To data, numerous studies have been carried out to investigate whether polymorphisms in APE1 had been extensively studied as the major AP endonuclease involved in the repair of AP sites through both its endonuclease and phosphodiesterase activities [APE1 may influence its capacity to repair DNA damage. Several lines of evidence support that the −656 T > G polymorphism plays a role in influencing the promoter activity of APE1. Using in vitro promoter assay, Lo et al. found that the −656 G allele had a significantly higher transcriptional activity than that of the −656 T allele and individuals with the −656 G allele were at a decreased risk for lung cancer [APE1 might offer protection against the development of lung cancer [APE1 and resulted in decreased transcriptional activity. Given the disparity between these functional studies, the mechanism underlying the regulation of APE1 by this polymorphism may be still required further investigation. Nevertheless, our meta-analysis showed that individuals with the variant genotypes (GG or GG/TG) of −656 T > G polymorphism were associated with a decreased cancer risk than those with the TT genotype. Although publication bias was observed in the recessive genetic model, our meta-analysis with or without trim-and-fill method did not draw different conclusions, suggesting that our results were statistically robust.Apurinic/apyrimidinic (AP) sites are common mutagenic and cytotoxic DNA lesions which caused by the loss of normal bases . APE1 hativities . Alteratg cancer . In theig cancer . Nearly g cancer . HoweverIn the subgroup analysis by cancer type, we found that individuals with the variant genotypes of −656 T > G polymorphism were associated with decreased lung cancer risk. However, the result should be interpreted with caution. Because, only four types of cancer  were included and the studies of each type were very limited, which may have insufficient power to reveal a reliable association. In the subgroup analysis by ethnicity, the results suggested that the association was more apparent among Asian populations. Several reasons may lead to the ethnic difference. First, differences in genetic background may cause the difference. Then, environmental or life style context which strongly vary between populations may play a role. Besides, other factors such as selection bias and different matching criteria may also result in the difference. In considering of the limited numbers of studies, in the future, large numbers of studies will be required to validate these associations.APE1 1349 T > G polymorphism and cancer risk.. Given the power of this meta-analysis which included 14,740 cases and 20,533 controls, a false-negative finding is unlikely. However, further studies may be still needed to investigate the interactions between APE1 polymorphism and environmental factors which play important role in carcinogenesis but were not assessed in the present meta-analysis due to lack of original data. Besides, more attention drew to the prospective −656 T > G polymorphism in the further studies may be helpful to reveal the role of APE1 in the etiology of cancer.Although it has been suggested the 1349 T > G polymorphism could influence the sensitivity to ionizing radiation , it doesAPE1 −656 T > G polymorphism. Secondly, lacking the original data of the included studies limited our study to further evaluate the potential interactions, since gene-environment and gene-gene interactions and even different polymorphic loci of the same gene may also modulate cancer risk. Besides, our results were based on unadjusted estimates, while a more precise analysis needs to be conducted if individual data such as age and sex are available. Nevertheless, advantages in our meta-analysis should also be acknowledged. First, a systematic review of the association of APE1 polymorphisms with cancer risk is statistically more powerful than any single study. Second, the studies included in our meta-analysis strictly and satisfactory met our selection criteria.Some limitations of our meta-analysis should be addressed. Firstly, the numbers of published studies collected in our analysis were not large enough for the comprehensive analysis, especially for the APE1 −656 T > G polymorphism, but not the APE1 1349 T > G polymorphism may contribute to genetic susceptibility to cancer. Further prospective studies with larger numbers of unbiased-matched homogeneous patients and well-matched controls are required to validate our results and to clarify the gene-gene and gene-environment interactions between APE1 polymorphisms and cancer risk.In conclusion, the results from this meta-analysis suggest that the The authors declare that they have no competing interests.BZ carries out the meta-analysis study and drafted the manuscript. HS participates in the design of the study and performs the statistical analysis. YS and KX collect and extract the data. XS and WM have been involved in revising the manuscript critically for important intellectual content. QS conceives of the study, and participates in its design and coordination and helps to draft the manuscript. All authors read and approve the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/11/521/prepub


Although meta-analyses are typically viewed as retrospective activities, they are increasingly being applied prospectively to provide up-to-date evidence on specific research questions. When meta-analyses are updated account should be taken of the possibility of false-positive findings due to repeated significance tests. We discuss the use of sequential methods for meta-analyses that incorporate random effects to allow for heterogeneity across studies. We propose a method that uses an approximate semi-Bayes procedure to update evidence on the among-study variance, starting with an informative prior distribution that might be based on findings from previous meta-analyses. We compare our methods with other approaches, including the traditional method of cumulative meta-analysis, in a simulation study and observe that it has Type I and Type II error rates close to the nominal level. We illustrate the method using an example in the treatment of bleeding peptic ulcers. Copyright © 2010 John Wiley & Sons, Ltd. Cochrane Database of Systematic Reviews The meta-analysis of results from multiple, similar studies is traditionally considered a retrospective activity. However, many meta-analyses are updated over time as new studies are undertaken or are identified from the previously hidden literature. In particular, systematic reviews in the et al. note that when there is no underlying effect the process of continually updating a meta-analysis using standard significance tests will lead eventually to a false-positive result If a meta-analysis is conducted repeatedly on the addition of new studies, without any allowance for multiple testing, the overall risk of a false-positive finding will increase with the number of meta-analyses performed. Indeed, Berkey et al. apply the law of the iterated logarithm that ‘penalizes’ the usual test statistic to account for multiple tests, equivalent to using a particular open ended boundary in the framework of Whitehead, although the approach does not control Type II error Whitehead I2, a measure of inconsistency across the observed studies' findings I2 might be used. Much of the present paper is devoted to the problem of accounting prospectively for heterogeneity, as it is the principal obstacle to the simple application of sequential trial methodology.Heterogeneity of effects is a key characteristic of many meta-analyses, and estimation of the among-study variance may create problems in the early stages of a sequential meta-analysis. The methods of Pogue have been adapted to account for heterogeneity in a (retrospective) cumulative meta-analysis scenario, by adjusting the desired sample size based on a function of In this paper we consider the use of frequentist sequential approaches to meta-analysis with the aim of controlling the risk of drawing incorrect conclusions. We take an approach based on ‘group-sequential’ trial methodology, with a stopping rule. A previously proposed method is reviewed and extended to better incorporate random effects. We compare six methods  through a simulation study and compare them in an example in the treatment of bleeding peptic ulcers. Our emphasis is on the determination of a straightforward approach that has good empirical properties, recognizing that even more formal methods are available through the use of dedicated software such as PEST yi, of a true effect θi with estimated variance vi, from study i, i = 1, …, k. The effect might be measured as a log odds ratio or a difference in means for a comparative study, or a simple mean or logit proportion for a single group study; numerous other options are possible. We make the usual assumption that the estimates and variances are uncorrelated, although this will not always be the case . Alternative approaches are available that do not rely on this assumption. For example, the raw data from the studies could be modelled, which may require individual participant data. Alternatively, all studies could be given equal weight Before introducing the sequential methods we review a general parametric approach to meta-analysis on which the sequential methods are based i across studies. A standard assumption is that this distribution is normal with mean µ and variance τ2. The variance describes the extent of heterogeneity. A simple estimate of µ is given by the weighted average i. An approximate variance of this estimate is given by 2 may be estimated simply by wi = 1/viand i across studies rather than on µ alone. This can be achieved using a predictive interval for the effect in a new study, which takes into account both the mean µ and variance τ2 of effects A random-effects analysis involves inference on the distribution of the θwi and the resulting calculations lead to a fixed-effect meta-analysis.If yi and vi are typically calculated from either Wald statistics or score statistics. For the former, i and yi = Zi/Viand vi = 1/Vi, where Zi is the efficient score for θi and Vi is Fisher's information, both evaluated at θi = 0. When sample sizes are large and θi is reasonably near 0, there will be little difference between the results from the two approaches.The estimates et al. introduce a desirable total that they refer to as the ‘optimal information size’, based on standard sample size calculations et al. adapt this to account for observed heterogeneity in a retrospective cumulative meta-analysis et al. yi and wi are calculated either from Wald statistics  or score statistics (as in Whitehead).To follow the progress of a meta-analysis over time we require a measure of the amount of relevant information contained within it. One approach is to use the total number of study participants tj studies have been included, j = 1, 2, …. A plot of Vj against the cumulative effect size, Zj, Vj) has slope equal to the simple random-effects meta-analytic estimate at the corresponding stage. Note that as information about τ2 changes over time, the individual study weights, A boundaries approach to sequential meta-analysis involves monitoring information using the sum of weights, 2 is assumed to be zero , this approach may be justified formally In the absence of heterogeneity, and when τet al. implement a double-triangular design Z, V) plane, with symmetric upper and lower boundaries at Z = ± H and a vertical boundary at V = Vmax reflecting the maximum amount of information to be collected  or that the mean effect is zero (if only the latter criterion is true). We refer to the values of Zj and Vj at this point as Zend and Vend. If the path stops with Vend>Vmax, then we extend the Christmas tree correction to Vend, and consider an effect to be present if |Zend|≥H′end.As soon as |j, study i has interim data yielding an estimate yij and weight 2 across currently available data), then we can calculate The methods we describe can be extended to address incremental information within studies as well as across studies. For instance, if at time Lj and Uj, for repeated confidence intervals after the inclusion of tj studies can be obtained by inverting the sequential design, yielding Zj≤ − H′j or if Zj≥H′j, is equivalent to stopping if the interval  excludes 0.The sequential procedure, described in the previous section in terms of monitoring boundaries, may also be presented using the repeated confidence intervals approach introduced by Jennison and Turnbull Lj, Uj) have the property that all of them, calculated up until V = Vmax, contain the true value of the parameter µ with probability 1 − α, where αis the chosen two-sided significance level. When the amount of information is less than Vmax, the confidence interval series has probability greater than 1 − α that the truth is contained in every interval. This means that at any time, an individual confidence interval has probability greater than 1 − α of containing the truth, and is therefore wider than a confidence interval from a conventional meta-analysis.The repeated confidence intervals  of the sequential meta-analysis will go backwards. The situation also has implications for the Christmas tree correction, and if Vj is smaller than Vj − 1 we set H′j = H, so that a correction is only made if the path goes forwards. Within a repeated confidence interval framework, a decrease from Vj − 1 to Vj yields a confidence interval wider than the previous one. We consider that the possibility that the confidence intervals can become wider (or that the path goes backwards)—and the actual widening of the confidence intervals—is a suitable reflection of the increased uncertainty resulting from identifying inconsistency among the studies.A key issue in sequential random-effects meta-analysis is the heterogeneity variance parameter, τ2  approximate Bayesian updating of uncertainty in τ2. The last two approaches are proposed to allow for prior information on τ2 to be incorporated in the early stages of the sequential meta-analysis. We refer to the sequential approaches using Bayesian inference for τ2 as ‘semi-Bayes’ to reflect the fact that inference on the parameter of interest (µ) is frequentist. We compare the five approaches, contrasting them with a naïve cumulative random-effects meta-analysis without correction for multiple looks, in a simulation study in Section 4.In Section 3 we consider five methods for addressing the heterogeneity variance in a sequential meta-analysis: (i) ignoring it ; (ii) implementing a standard random-effects method (using the DerSimonian–Laird estimate); (iii) accounting for uncertainty in τyi and wi are as defined in Section 2.1. As with all methods described below, we apply the confidence interval formulae in Equation  path towards the boundary, since the uncertainty is reflected in a smaller value for V. Early repeated confidence intervals are likely to be wide.As the new weights, 2 in the presence of a small number of studies is to draw on external evidence about its likely value. An informative prior distribution may contribute heavily to producing a realistic estimate in the early stages of a sequential meta-analysis. We may therefore expect the estimates of τ2 in the meta-analysis to be less erratic, and the path to proceed forward towards the boundaries. We propose to use Bayes' theorem to update the value of τ2, and to insert this into the frequentist sequential approach outlined above.A natural approach to making inferences about τ2 may be used in subsequent updates of the sequential meta-analysis.A suitable prior distribution may be developed following the ideas of Higgins and Whitehead p(τ2) represent the prior distribution for τ2, which here will be considered to be an inverse gamma distribution with shape parameter η and scale parameter λ. At the jth meta-analysis the likelihood function for the data L is a product of normal distributions with means equal to µand variances equal to (vi + τ2). The joint posterior distribution of µand τ2 at this meta-analysis, assuming an independent prior distribution, p(µ), for µ, is given by 2 would normally be performed by integration or simulation. However, µ is unknown and in our frequentist analysis is not associated with a probability distribution, so cannot be integrated out. A solution is to replace µ by its estimate from the (j − 1)th meta-analysis. This makes τ2 the only unknown parameter in the posterior formula, and a simple one-dimensional numerical integration can be performed to obtain its posterior mean, j = 1) contains several studies (i.e. t1>1), then a suitable value for t1 − 1) studies. If the first meta-analysis contains exactly two studies (t1 = 2) we would set y1.Let 2 in the early stages, producing an analysis that is less likely to yield a false positive result. As information accumulates, and the posterior distribution of τ2 develops a stronger peak, the mean will move towards the median and the mode.We opt for the mean rather than the median or mode since not only is it straightforward to calculate, but it will tend to overestimate τ2 by applying the ideas of Biggerstaff and Tweedie. Specifically, the integrations in the calculation of For a random-effects sequential meta-analysis, we take yi = θi and vi = 0. Then the inverse gamma prior distribution for τ2 is the conjugate prior distribution for the unknown variance of a normal distribution, and the posterior distribution for τ2 is then an inverse gamma posterior distribution with parameters (η + (tj/2)) and tj. We may estimate this variance using the DerSimonian–Laird estimate 2 is given by 2 and the usual estimate from the studies in the meta-analysis. As the number of studies increases, the effect of the prior information decreases.We now propose a simpler approach that avoids the numerical integration. Suppose the true effects in the studies to date were known exactly, so that For a random-effects sequential meta-analysis, our approximate semi-Bayes approach takes yi≈θi and vi≈0), and to overestimate τ2 when the studies are small, due to the overdispersion in the distribution of the estimates, yi, compared with the distribution of the true effects, θi.In practice, the approximations here might be expected to yield adequate results when the studies are large  distribution. The mean within-study variance of the effect estimates, yi, across all studies was taken to be σ2 = t/Vmax, ensuring that there would be approximately t studies in a sequential fixed-effect meta-analysis if the vertical boundary was crossed at Vmax and τ2 = 0. The within-study variance of the effect estimate for each study was sampled from a uniform distribution in the range . This ensures that the studies have different degrees of precisions, reflecting a realistically wide range of study sizes within each meta-analysis. As an approximate interpretation, for the case of one-group studies estimating means with a between-person standard deviation equal to 4 in every study, this ensures that, for t = 5, sample sizes would be distributed between 81 and 567 rather than all being equal to 142. The prior distribution for τ2 for the semi-Bayes methods was an inverse gamma distribution with parameters η = 1.5 and λ = 0.08. This corresponds to a prior estimate for τ2 of 0.16. This is less than the maximum simulated heterogeneity, and, if the true effect size were 0.5, allows for some studies to have negative effect sizes. Taking η = 1.5 is equivalent to having prior information with the weight of one study, so the prior estimate contributes less than 10 per cent towards the posterior estimate of heterogeneity after ten studies have been included in the meta-analysis.Within each simulated sequential meta-analysis the true effect size from each study, θVj exceeds Vmax; or otherwise to continue. The simulation was conducted in R (the program is available upon request).For each of the 36 scenarios 5000 sequential meta-analyses were simulated and analysed using each of the six approaches, using the procedures described in Section 2. The stopping rule for the naive cumulative meta-analysis method was to stop (conclude an effect exists) if the naive 95 per cent confidence interval excludes 0; or to stop (conclude no effect) if Zend and Vend) contained the true effect size µ; (v) estimates of the heterogeneity variance, τ2, at the time when a stopping boundary is crossed and (vi) mean value of Vend.From the simulations, for each approach, we recorded (i) the mean number of studies taken to reach a stopping boundary of the design; (ii) the proportion of simulations stopping having crossed the upper boundary ; (iii) the percentage of simulations in which all calculated two-sided 95 per cent confidence intervals contained the true effect size µ ; (iv) the percentage of simulations in which the confidence interval at the time when a stopping boundary is crossed , t = 10  and t = 20 . The results of the simulations should be interpreted in comparison with the fixed-effect sequential method as well as in isolation. The fixed-effect sequential method should in theory have good properties in the absence of heterogeneity. However, the results do not agree with the expectations due to a combination of (i) the inadequacy of the Christmas tree correction; (ii) our random sampling of within-study variances from a uniform distribution, which means that we cannot guarantee to hit the Vmax boundary with the tth study and (iii) the ad hoc decision to consider an effect present when Vend>Vmax and |Zend|≥H′end, a region in which sequential theory does not formally apply. The third issue is expected to increase the probability of declaring that an effect exists, with a larger increase the further away Vend is from Vmax. When µ = 0, the vertical boundary will be crossed in the majority of simulations, and it can be seen from 2 = 0) with approximately five studies to reach Vmax that the Type I error rate is 0.039 instead of 0.025. When there are more and smaller studies, for example in Vend moves closer to Vmax and it can be seen that the Type I error rate reduces to 0.028. In addition, the probability that all confidence intervals up to and including the one at Vend will contain the true effect size will only be equal to 0.95 if Vend = Vmax. If Vend<Vmax, this probability will be greater than 0.95, and if Vend>Vmax this probability will be less than 0.95. The closer that Vend is to Vmax, the closer will be the probability to 0.95. When µ = 0 and τ2 = 0, this probability is equal to 0.922 . From  = 23.07 .2 are used for the semi-Bayes methods: an IG  has a prior mean of 0.16 and an IG  has a prior mean of 2. These prior means are respectively much smaller and much larger than the DerSimonian–Laird estimate based on all of the trials. Zend/Vend) and heterogeneity, and the last repeated confidence interval. If the plan for this sequential meta-analysis had been to follow the formal stopping rule of the O'Brien and Fleming design, then it would be more appropriate to present a ‘final’ analysis which correctly adjusts for earlier looks. The results from conducting this ‘final’ analyses using the PEST package Zend/Vendestimate, and provides an appropriate P value and confidence interval. The confidence interval is narrower than the last repeated confidence interval, because the repeated confidence intervals have had no effect on the decision to conduct further studies.The sequential random-effects meta-analysis and the approximate semi-Bayes and semi-Bayes methods are applied and compared with a sequential fixed-effect meta-analysis . To asse2 than is apparent in the dataset, as with the IG  prior, the semi-Bayes method stops after a smaller number of studies (nine) although the approximate semi-Bayes method carries on for the same 11 studies as for the sequential random-effects method. The earlier stopping of the semi-Bayes IG  method results in smaller estimates of treatment effect and heterogeneity. The use of an IG  prior with its large prior estimate of heterogeneity causes the approximate semi-Bayes and semi-Bayes methods to stop after 15 trials. The estimates of treatment effect and heterogeneity from the sequential approaches are smaller than those from the overall random effects analysis of 23 studies. This is mainly due to the seventeenth trial, which has a large, beneficial effect of treatment, increasing the heterogeneity.These results show that a sequential fixed-effect meta-analysis, which ignores the heterogeneity, stops after four studies see also , leadingyi and vi. Indeed, a commonly applied test for association between the estimates and standard errors produces P = 0.005 P = 0.44 In our example, we have assumed that a sequential approach to the meta-analysis had been planned before the trials had reported their results. The example is instructive in illustrating some of the limitations of applying our generic statistical methods prospectively in a meta-analysis situation. We have remarked already that 2 × 2 contingency tables may be better analysed using binomial likelihoods. This would have avoided the need to make continuity corrections by adding 0.5 s to cell counts. Furthermore, the use of maximum likelihood estimates of odds ratios also creates a correlation between the Vmax is reached, then the recommendation can be made that no further research is required to inform the question addressed by the meta-analysis.This paper has considered the use of sequential methods for meta-analyses that incorporate random effects to allow for heterogeneity between studies. Our emphasis has been on determining a straightforward approach that has good empirical properties. We have developed and compared five methods, including a direct extension of the standard random-effects meta-analysis method, an extension of the approach of Biggerstaff and Tweedie, a semi-Bayes method involving Bayesian updating of the heterogeneity variance, and an approximation to the semi-Bayes method. The choice of an O'Brien and Fleming sequential design with a Christmas tree correction for discrete monitoring leads to a simple procedure for calculating repeated confidence intervals, which enables us to present sequential meta-analyses using forest plots, the conventional way of illustrating meta-analysis results. When either a repeated confidence interval excludes zero or Some issues are worthy of discussion, including philosophical, practical, and technical considerations. First, should sequential methods be applied to meta-analysis? Second, if sequential methods are to be used, what methods should be implemented in practice? Third, what technical problems remain that require further research?The use of sequential methods for meta-analysis is contentious. Whereas the steering committee for a prospective study has control over the recruitment of participants, researchers undertaking meta-analyses may have no direct control over whether further studies are performed. Whether sequential methods have a role for updating of meta-analyses that are unconnected with primary researchers (such as those in Cochrane reviews) remains an open question. Chalmers and Lau Vmax can be determined. However, there are many considerations in the design of a new study, and it is rare for them to be designed primarily around an updated meta-analysis. Related issues are discussed by Sutton et al. Meta-analyses are now well established as part of the research process, and many have argued for the need for primary studies to both be informed by meta-analyses of existing evidence, and be rounded off with an updated meta-analysis including the newly generated evidence. Thus primary research is increasingly viewed as part of a wider sequential process, and the methods described here may have application in this context. For instance, there may well be more enthusiasm for conducting a new study if the effect estimate is close to an upper or lower stopping boundary in a scheme such as One area of potential application of sequential methods is the monitoring of adverse effects of pharmacological interventions. However, as we remark in Section 5, the particular general parametric approach we describe does not lend itself well to rare events. Furthermore, it is not clear that strict control over false-positive findings is important in this context, since a small, non-statistically significant, signal should still be investigated when the adverse effect is major. More research is required to explore appropriate methods for this issue.2, but an explicit stopping rule does not follow from such an approach.If sequential methods are to be implemented, can the methods we have described be recommended? Certainly we would recommend against a fixed-effect approach. However, in a random-effects framework, a key problem in the early stages of a sequential meta-analysis is the poor estimation of the heterogeneity variance. We argue that prior information is therefore necessary. Of our two semi-Bayes methods, the approximate semi-Bayes approach is straightforward to implement, and this is the one we recommend from among those we investigated. From the simulation results, it appears that the method has reasonable false-positive and coverage properties, although these are likely to be dependent on the choice of prior distribution for the among-study variation. Our application demonstrated that the early stages of the sequential meta-analysis can be affected by the prior distribution. Further empirical research is needed to characterize the degree of heterogeneity that can be anticipated in a meta-analysis with particular clinical and methodological features, so that realistic informative prior distributions can be formulated. An alternative approach would be a sensitivity analysis in which the sequential meta-analysis is repeated for a variety of plausible values of τet al. Vmax is exceeded. When this last repeated confidence interval is calculated, there is a probability of slightly less than 1 − α that the truth is contained in every interval. The reduction in the probability depends on the increase in V from Vmax, and will be greater if only a few sequential steps have been conducted. A better understanding of this issue would be useful. Finally, the advantage of using the Christmas tree correction for discrete monitoring with the O'Brien and Fleming design is that this leads to a simple procedure for calculating the repeated confidence intervals, but there are also some disadvantages to this approach. First, Stallard and Facey Vj) and previous (Vj − 1) meta-analysis update. If (Vj − Vj − 1) is less than 0, there is no Christmas tree correction. However, if at the next update (Vj + 1 − Vj) is greater than 0, there will be a Christmas tree correction. However, Vj + 1 may be less than Vj − 1. Further work is needed to assess the effect that this might have on the validity of the results.Our recommendations are interim, since more work is required in this area. We have commented on methods based on the law of the iterated logarithm proposed by Lan 2 to be incorporated. The entire meta-analysis could alternatively be undertaken within a fully Bayesian framework. The meta-analyst would then calculate a new credibility interval for every meta-analysis update. The credibility interval, unlike the frequentist repeated confidence interval, is not corrected for multiple looks. The posterior distributions of µ and τ2 from the first meta-analysis would become the prior distribution for the second and so on. If τ2 increases during the course of the sequential meta-analysis, the credibility interval may become wider, but this causes no problem. Bayesian stopping rules for individual clinical trials have been proposed by various authors (see, for example We have considered a frequentist sequential approach to meta-analysis, proposing semi-Bayes approaches to allow prior information about τ


VEGF gene polymorphisms and lung cancer risk has been extensively studied in recent years, but currently available results remain controversial or ambiguous. The aim of this meta-analysis is to investigate the associations between four common VEGF polymorphisms  and lung cancer risk.Vascular endothelial growth factor (VEGF) is one of the key initiators and regulators of angiogenesis and it plays a vital role in the onset and development of malignancy. The association between VEGF polymorphisms and lung cancer risk. Crude odds ratios (ORs) with 95% confidence intervals (CIs) were used to evaluate the strength of this association.A comprehensive search was conducted to identify all eligible studies to estimate the association between VEGF −2578C>A polymorphism is capable of increasing lung cancer susceptibility, especially among smokers and lung squamous cell carcinoma (SCC) patients. Additionally, for +936C>T polymorphism, increased lung cancer susceptibility was only observed among lung adenocarcinoma patients. In contrast, VEGF −460C>T polymorphism may be a protective factor among nonsmokers and SCC patients. Nevertheless, we did not find any association between +405C>G polymorphism and lung cancer risk, even when the groups were stratified by ethnicity, smoking status or histological type.A total of 14 published case-control studies with 4,664 cases and 4,571 control subjects were identified. Our meta-analysis provides strong evidence that This meta-analysis recommends more investigations into the relationship between −2578C>A and −460C>T lung cancer risks. More detailed and well-designed studies should be conducted to identify the causal variants and the underlying mechanisms of the possible associations. Therefore, we performed a meta-analysis of all eligible case-control or cohort studies to investigate whether these functional VEGF polymorphisms are associated with any increased risk of lung cancer and whether the associations are modulated by smoking status, histological type or other risk factors. We hope our meta-analysis can potentially be important in early lung cancer identification and become part of the therapeutic strategies in combating lung cancer.The VEGF gene polymorphisms and lung cancer risk. As the main search criteria, we used combinations of the following terms: "VEGF", "vascular endothelial growth factor A", "vascular permeability factor", "vascular endothelial growth factor", "lung neoplasms", "pulmonary neoplasms", "bronchial neoplasms", "lung cancer", "bronchial neoplasm", "genetic polymorphism", "single nucleotide polymorphism", "SNP", "mutant", "gene variation". We also reviewed the reference lists of articles retrieved to identify relevant publications.Relevant papers for this meta-analysis were systematically identified through literature searches on PubMed, Embase, Web of science and Chinese National Knowledge Infrastructure (CNKI), and Chinese Biomedical Literature Database (CBM) of publications published up to March 9, 2013 relating to VEGF gene and lung cancer risk; (b) the diagnosis of lung cancer patients was confirmed pathologically and controls were confirmed as cancer-free patients; (c) inclusion of sufficient data on the size of the sample, odds ratio (OR), and 95% confidence interval (CI) and (d) articles were published in the English or Chinese language.Our meta-analysis included genetic association studies fulfilling the following inclusion criteria: (a) a case-control, cohort or cross-sectional study must evaluated at least one of four polymorphisms of Studies were excluded when they represented duplicates of previous publications, or were meta-analyses, case report, letters, reviews or editorial articles. Studies investigating the progression, severity, phenotype modification, response to treatment, or survival were also excluded. Additionally, when data was included in multiple studies using the same case series, either the study with the largest sample size or most recent publication was selected. Finally, family-based studies were excluded because of different design settings. Any disagreements on study inclusion were resolved through discussions between the authors. To ensure the rigour of the current meta-analysis, it was designed and reported according to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement. The relevant checklist is shown in All data from the included studies were extracted independently by two investigators, using a piloted data standardized form : the first author’s surname, year of publication, country of origin, published language, gender of study individuals and ethnic subgroups, study design, number of subjects, smoking status, histological types of lung cancer, SNP genotyping methods, genotyping method and detected sample, allele and genotype frequencies, and evidence of Hardy-Weinberg equilibrium (HWE) in controls. In addition, we also compared key study characteristics such as location, study time and authorship to determine the existence of multiple publications from the same study.Two authors independently assessed the quality of the published articles according to the modified STROBE quality score systems VEGF gene polymorphisms and lung cancer risk under five genetic models: allele, dominant, recessive, homozygous, and heterozygous models. The deviation of frequency from those expected under Hardy-Weinberg equilibrium (HWE) was assessed by Chi-squared goodness of fit tests in controls. We explored inter-study variation through prespecified subgrouping of studies according to ethnicity , gender , smoking status , and histological type of lung cancer , and small cell carcinoma (SCLC), where applicable. The statistical significance of the pooled OR was assessed with a Z test. Between-study variation and heterogeneity were estimated using Cochran’s Q-statistic, with P<0.05 as a cutoff for statistically significant heterogeneity Crude ORs together with their corresponding 95% CIs were used to calculate and assess the strength of association between 2I test (ranges from 0 to 100%), which represents the proportion of inter-study variability that can be attributed to heterogeneity rather than to chance Mantel-Haenszel method) was used, except when a significant Q-test (P<0.05) or 2I>50% indicated the existence of heterogeneity among studies; otherwise, the random effects model (DerSimonian-Laird method) was applied for meta-analysis. In order to ensure the reliability of our results, sensitivity analysis was performed by omitting individual studies. Begger’s funnel plots were used to detect publication bias. In addition, Egger’s linear regression test, which measures funnel plot asymmetry via a natural logarithm scale of OR, was also used to evaluate publication bias We also quantified the effect of heterogeneity with the VEGF gene, −2578C>A, −460C>T, +936C>T, and +405C>G. The flow diagram of the selection of studies and specific reasons for exclusion from the meta-analysis are shown in VEGF SNPs in 4,664 unrelated lung cancer cases and 4,571 unrelated controls from 14 case-control studies. In the eligible studies, there were 12 studies of subjects of Asian descent and only two studies of subjects of Caucasian descent. All included studies extracted DNA from peripheral blood and the VEGF polymorphisms were determined by classic PCR-RFLP in 12 studies, by TaqMan in 1 study, and by PIRA-PCR in another study. SNP genotypes were tested for departures from HWE for controls and all SNPs were in HWE. The qualities of the included studies were moderately high, with a STROBE score of greater than 20. The selected study characteristics were summarized in VEGF −2578C>A, −460C>T, +936C>T, and +405C>G polymorphisms and lung cancer risk are presented in Our initial literature search yielded 546 reports, which included 13 population-based P = 0.003; OR = 1.79, 95%CI = 1.30–2.46, P<0.001). We also performed a stratified analysis based on the geographic region of the studies; the results were persistent in both Chinese  and India populations  , while not in nonsmokers    . Additioarisons) . FurtherP<0.001) .VEGF −460C>T polymorphism was investigated in six studies with a total of 3,167 cases and 2,769 controls. There was no evidence of association between −460C>T polymorphism and lung cancer risk . Again, lung cancer cases and controls did not significantly differ in the subgroup analyses based on ethnicity (data not shown)  , but not in adenocarcinoma and SCLC patients  (t shown) . HoweverP<0.001) . Additioarisons) .P = 0.301), even when the groups were stratified by ethnicity , gender , or smoking status . Nevertheless, in the subgroup analysis on histological subtype, increased lung cancer susceptibility was shown among the adenocarcinoma subgroup   .P = 0.840; dominant model: OR = 1.23, 95%CI: 0.60–2.50, P = 0.573; recessive model: OR = 0.815, 95%CI: 0.62–1.05, P = 0.108; homozygous model: OR = 0.92, 95%CI: 0.62–1.37, P = 0.689; heterozygous model: OR = 0.70, 95%CI: 0.46–1.07, P = 0.099). Likewise, stratified analyses were also conducted according to ethnicity, smoking status and histological type of cancer. Unfortunately, there was no statistical difference in genotype distributions between cases and controls, and overall and different subgroups .A total of six studies involving 3,035 cases and 2,880 controls were included in the pooled analysis. The overall data did not show a marked association between +405C>G polymorphism and lung cancer risk in any genetic model, indicating that individuals with G allele might not have an increased lung cancer risk compared with those who carry wild-type C allele , which indicates that our studies were statistically accurate.Sensitivity analysis was performed to assess the influence of each study on the pooled ORs by omitting individual studies. The analysis results suggested that no individual study significantly altered the pooled ORs in VEGF −2578C>A (A), −460C>T (B), +936C>T (C), and +405C>G (D) polymorphisms .Begger's funnel plot and Egger's linear regression test were performed on the metadata to assess publication bias of the individual studies. The shapes of the funnel plots did not reveal any evidence of obvious asymmetry in orphisms . Egger'sVEGF, as a predominant angiogenic factor in human cancers, plays a vital role in the carcinogenesis pathway, which has been proved to be a key step in tumor occurrence, progression and prognosis VEGF gene have been confirmed to be correlated with high levels of VEGF protein in cancer cells and high tumor angiogenic activity, and they also contribute to the susceptibility and severity of cancer, including lung cancer VEGF gene polymorphisms develop tumors with low VEGF expression and poor vascularization VEGF and risk of lung cancer have been extensively investigated, obtaining conflicting results. Therefore, we employed a meta-analysis to explore a more precise evaluation for the associations. To our knowledge, this is the first meta-analysis on this topic.Evidence from preclinical and clinical studies shows that VEGF −2578C>A, −460C>T, +936C>T, and +405C>G polymorphisms and lung cancer risk. According to our pooled analysis, −2578C>A polymorphism may have a correlation with increased lung cancer risk. This finding may be biologically plausible since Koukourakis et al. observed that −2578CC was associated with lower VEGF expression and lower vascular density in lung cancer tissues compared to the −2578C C/A The present meta-analysis, including 4,664 cases and 4,571 controls from 14 published case-control studies, explored the association between VEGF −460C>T polymorphism, the overall data did not show a marked association of this polymorphism with lung cancer risk in any genetic model, even in the subgroup analyses according to ethnicity. However, when stratified analysis by smoking status and histological type were performed, a lower prevalence of −460T allele was observed among nonsmokers, lung adenocarcinoma cases, and SCC cases. Some clinical evidence suggests that cigarette smoking may stimulate both angiogenesis and VEGF expression, which exacerbates the rapid cancer progression effect of angiogenesis VEGF activate multiple effects in lung cancer. For VEGF +936C>T, +405C>G polymorphisms, we found no overall association between these two polymorphisms or its interaction with smoking on lung cancer risk in any genetic model. When stratified analyses were conducted according to ethnicity and histological types of cancer, increased lung cancer susceptibility was only observed among the adenocarcinoma subgroup for +936C>T polymorphism, while there was no statistical difference in genotype distributions between cases and controls for any different subgroups for +405C>G polymorphism. Actually, there exist conflicting reports in some literatures regarding the exact function of the +405G/C polymorphism. Some clinical studies suggested that +405C allele has been associated with lower VEGF production, while some groups reported higher VEGF levels or even no association with +405C/C genotype As for VEGF, and our meta-analysis emphasized that elucidating the pathogenesis of lung cancer would demand an investigation into the association for many gene variants that may constitute distinct pathophysiological pathways. Third, we identified two studies from Caucasian populations and obtained no data from African populations, thus the two racial groups need to be further studied in the future. Therefore, the results should ideally be confirmed in further studies to strengthen the conclusions. Aside from the limitations listed above, our meta-analysis still has some strength. To the best of our knowledge, this is the first meta-analysis on the relationship between VEGF gene polymorphisms and lung cancer. We also explored inter-study variations by prespecified subgrouping of studies according to ethnicity, smoking status, gender, and histological type among cases. Furthermore, although this meta-analysis does not accommodate all previously published data, they are limited compared to the evidence that we generated.The current meta-analysis has several limitations that should be noted. First, the sample size in the present study was relatively small, so small, but potential, genetic effects may not be detectable. A small sample size may not have enough statistical power to explore the real association, especially in subgroup analysis. Additionally, as with other complex traits, lung cancer risk may also be modulated by several other genetic markers beyond VEGF −2578C>A polymorphism is capable of increasing lung cancer susceptibility, especially among smokers and lung SCC patients. Additionally, for +936C>T polymorphism, increased lung cancer susceptibility was only observed among lung adenocarcinoma patients. In contrast, VEGF −460C>T polymorphism may be a protective factor among nonsmokers, lung adenocarcinoma and SCC patients. However, we did not find any association between +405C>G polymorphism and lung cancer risk, even when the groups were stratified by ethnicity, smoking status or histological type. More detailed and well-designed studies with larger population and different ethnicities are needed to further evaluate these associations.In conclusion, this meta-analysis provides strong evidence that Supplement S1PRISMA Checklist.(DOC)Click here for additional data file.Supplement S2Modified STROBE quality score systems.(DOC)Click here for additional data file.