APE1/Ref-1) is implicated in tumor development and progression. Recently, the APE1/Ref-1 promoter -141T/G variant (rs1760944) has been reported to be associated with lung cancer risk. Given the importance of APE1/Ref-1 in both DNA repair and redox activity, we speculate that the -141T/G polymorphism may confer individual susceptibility to gliomas or its subtypes.The human apurinic/apyrimidinic endonuclease 1/Redox effector factor-1  and 824 cancer-free controls from eastern China. Genotyping was performed with Sequenom MassARRAY iPLEX platform by use of allele-specific MALDI-TOF mass spectrometry assay. We estimated odds ratios (ORs) and 95% confidence intervals (95% CIs) using unconditional logistic regression. A test of trend was calculated using the genotype as an ordinal variable in the regression model. For each statistically significant association identified, we estimated the false positive reporting probability (FPRP). FPRP values less than 0.2 were consider to indicate robust associations.The APE1/Ref-1 promoter -141T/G polymorphism and glioma risk was not observed. However, the stratified analysis by histology revealed the variant allele G significantly decreased glioblastoma risk . Individuals with the homozygous -141GG genotype exhibited 46% reduced risk of glioblastoma , compared with the TT homozygote. This result remained robust given the prior probabilities of 25% (FPRP = 0.052) and 10% (FPRP = 0.140), but not with a prior probability of 1% (FPRP = 0.643). The P-associated with the trend test was 0.014.The significant association between the APE1/Ref-1 promoter may modulate risk of glioblastoma, but not for other histological gliomas. Larger studies with more APE1 polymorphisms are required to validate these preliminary findings.Our results suggest that a specific genetic variant located in the  Gliomas present a group of primary brain tumors that originate from glia (from the Greek for "glue"), accounting for more than 40% of the newly diagnosed brain tumors . In accoAPE1/Ref-1) plays a central role in the base excision repair pathway for the repair of DNA damage generated by alkylation, oxidatation and IR [APE1/Ref-1 hydrolyzes the phosphodiester bond 5' to abasic sites, leaving a normal 3'-hydroxyl group and 5'-deoxyribose phosphate terminus for DNA repair synthesis [APE1/Ref-1 has 3'-repir diesterase or phosphatase activity [APE1/Ref-1 also possesses a 3'-5' exonuclease activity, reported to play a role in the excision of deoxyribonucleoside analogs from DNA [APE1Ref-1 also functions as a redox activator of numerous cellular transcription factors, such as AP-1, NF-кB, p53, Egr-1, c-Myb, HIF-1α, HLF, and Pax-8, that are thought to be important in carcinogenesis [APE1/Ref-1 in mice causes postimplanatation embryonic lethality [APE1/Ref-1 - knockout cell lines has been so far unsuccessful, clearly indicating that the importance of its function to cell survival and propagation.Exposure to IR can result in various types of DNA damage that cause cells to turn cancerous. Cells respond to this damage through the activation of various DNA repair pathways, and each pathway involves numerous molecules. The human apurinic/apyrimidinic endonuclease 1/Redox effector factor-1  at codon 148 of exon 5 of the APE1/Ref-1 gene . Although the APE1/Ref-1Asp148Glu polymorphism doses not result in reduced endonuclease activity [APE1/Ref-1 Glu allele alone was associated with an increased risk of breast caner [APE1/Ref-1 polymorphisms have also been evaluated for altered cancer susceptibility. One study reported that the non-synonymous SNP Ile64Val (rs2307486) was associated with a decreased risk of non-small cell lung cancer (NSCLC) [APE1/Ref-1 gene and other DNA repair genes may interact to have joint effects on risk of certain cancers [The Human activity . Geneticst caner  and gastst caner , and witst caner  and gliost caner , whereasst caner , pancreast caner  and gliost caner . For lunst caner , but no st caner -29. Othe (NSCLC) . Another (NSCLC)  or lung  (NSCLC) . Moreove cancers -21,23-28APE1/Ref-1 gene. The role of APE1/Ref-1 promoter polymorphisms in cancer risk was little studied. Recently, the -141T/G variant (rs1760944) in the APE1/Ref-1 promoter has been reported to be associated with a decreased risk of lung cancer in Chinese populations [APE1/Ref-1 in both DNA repair and redox activity, we speculate that APE1/Ref-1 -141T/G variant may also confer individual susceptibility to gliomas or its certain subtypes. To test this hypothesis, the APE1 promoter polymorphism was investigated in a hospital-based case-control study with 766 glioma patients and 824 cancer-free controls from eastern China. Odds ratios (OR) were determined for all study patients and for those with histological subtypes of glioblastoma (241 cases), astrocytomas except for glioblastoma (284 cases), and other gliomas (241 cases).Notably, these previous studies mainly focused on the non-synonymous SNPs of the ulations ,29. In vThe current study included 766 glioma patients and 824 healthy cancer-free controls. According to histological diagnosis, the glioma patients were stratified into three subgroups: 241 glioblastomas, 284 astrocytomas except for glioblastoma and 241 other gliomas. The subject enrollment commenced in October 2004 and is ongoing. All newly diagnosed glioma cases were inpatients for tumor resection at the Neurosurgery Department of Huashan Hospital of Fudan University, Shanghai, China. This hospital is the biggest center of neurosurgery in the east of China . HistopaEach eligible subject was face-to-face interviewed by a trained personnel who was blinded to the case-control status with a shortened structured questionnaire, which was adapted from the original version provided by Dr. Melissa Bondy for brain study in the Department of Epidemiology at M.D. Anderson Cancer center, to obtain data on demographic factors, smoking status, radiation exposure, family history of cancer and other factors. Each subject provided the informed consent and donated 3-5 ml of peripheral blood. As a result, DNA samples and questionnaires were available from 766 case and 824 control subjects representing an 85.7% and 79.2% of all eligible case and control subjects, respectively. The research protocol was approved by the Ethics Committee for Human Subject Research of Fudan University.Genomic DNA was extracted from whole-blood samples using the Qiagen Blood Kit , according to the manufacturer's instructions. After extraction, genomic DNA was diluted to a final concentration of 15-20 ng/μl for the genotyping assays. Polymorphism spanning fragments were amplified by the polymerase chain reaction (PCR) and performed genotyping with Sequenom MassARRAY iPLEX platform by use of allele-specific MALDI-TOF mass spectrometry assay . Primers2 test (df = 2). The Hardy-Weinberg Equilibrium was determined using the standard χ2 test (df = 1) to compare the observed frequency with the expected frequency in controls. To estimate the association between the APE1/Ref-1 -141T/G variant and risk of gliomas, odds ratios and 95% confidence intervals (95% CI) were calculated using unconditional logistic regression with adjustment for age and gender. The homozygosity with the more frequent allele among controls was set as the reference group. A test of trend was calculated by treating the three genotypes  as ordinal variables in the regression models. Furthermore, we performed stratified analyses for the APE1/Ref-1 -141T/G polymorphism by age (two groups: < 18 years and > 18 years), gender, smoking status, family history of cancer and histologic type. The SPSS 15.0 software  was used for all statistical analyses. All P values presented were two sided, and a level of P < 0.05 was considered statistically significant.Differences in demographic variables, smoking status, family history of cancer, and grouped allele and genotype frequencies between cases and controls were compared by the χAPE1/Ref-1 -141T/G variant, we considered that the genetic variant has been previously shown to influence expression of the APE1/Ref-1 gene [APE1/Ref-1 promoter polymorphism and glioma risk. Hence, we selected more conservative prior probabilities, 1% and 10%. In accordance with Wacholder et al [To further assess the probability of a spurious association due to multiple testing, we calculate the false-positive report probability (FPRP) for a SNP from an estimated OR and 95% CIs using the methodology described by Wacholder et al . To assif-1 gene ,29 and hf-1 gene ,29. In ter et al , a standP = 0.353, 0.919, and 0.535, respectively). The mean age of cases (42.4 years) at diagnosis was slightly older than the reference age for controls (41.5 years). 59.1% cases and 59.9% controls were male. As shown in the previous study, the cases were more likely to report a history of cancer in first-degree relatives than controls . Among 766 glioma cases, 241 were glioblastoma, 284 astrocytomas except for glioblastoma and 241 other gliomas (Table The details of selected characteristics of 766 glioma cases ad 824 control subjects have been shown in Table P = 0.262). Because there was no difference in allele and genotype frequency distribution between two control groups, "annual check-up subjects" and "trauma outpatients" (data not shown), we combined these two control groups for subsequent analyses.The genotyping success rate was 98.5%. The observed G allele frequency for rs1760944 T/G SNP in controls was 45.0%, which was similar to that reported in HapMap database for Chinese Han population. Genotype frequencies of rs1760944 T/G SNP among the controls did not differ significantly from those expected under HWE . In logistic regression analysis, neither the homozygous -141GG nor heterozygous TG genotype was associated with risk of gliomas in overall population. Further analysis showed that rs1760944 T/G genotypes were not associated with gender, age, smoking status, or family history of cancer (Table Allele and genotype frequencies and associated ORs (95% CI) for glioma cases and controls are presented in Table APE1/Ref-1 rs1760944 by glioma histology. As shown in Table P = 0.032 and 0.041, respectively). Overall, the variant G allele was associated with a decreased risk of glioblastoma compared with the T allele . Individuals with the homozygous -141GG genotype exhibited 46% reduced risk of glioblastoma , compared with the TT homozygote. According to the criteria suggested by Wacholder et al [P-associated with the trend test was 0.014 (P for trend = 0.014). However, no significant association of rs1760944 with risk of low-grade astrocytomas (astrocytomas except for glioblastoma) and other gliomas was found in this study. As the number of glioblastoma subjects was not large enough for stratification analysis, we did not further evaluate the association between the rs1760944 T/G and risk for glioblastoma by age, gender, smoking status and family history of cancer.Next, we performed the stratified analysis for http://www.psycho.uni-duesseldorf.de/abteilungen/aap/gpower3). It showed that with a sample size 1052 (236 glioblastoma cases and 816 controls), the study had > 90% power in detecting associations of rs1760944 polymorphism with risk of glioblastoma, at a significance level of 0.05 (df = 2).We also performed post-hoc power calculation based on the observed variant allele frequencies in both control and case groups, using post hoc power analyses of G*Power3  in the APE1/Ref-1 -141T/G polymorphism and lung cancer risk [APE1/Ref-1 -141TG or GG genotype was associated with a significantly decreased risk compared with TT homozygote in a hospital-based case-control study including 730 lung cancer cases and 730 cancer-free controls in Taiwan. Furthermore, the decreased risk was observed among the heavy smokers, but not among the light or never smokers, which reflected a gene-smoking interaction in the development of lung cancer. In two studies comprising a total of 1072 lung cancer patients and 1064 cancer-free controls in Chinese populations, Lu et al. [APE1/Ref-1 -141GG genotype against lung cancer risk, particularly among old subjects, current smokers, and subjects with a positive family history of cancer in their first-degree relatives. These data suggested that these variables may modify the effect of the -141T/G polymorphism on lung cancer. However, such gene-environment interactions were not observed in our study. Given that smoking is not associated with gliomas [Recently, two independent research groups have reported the association between the cer risk ,29. Lo ecer risk  found thAPE1 may be a possible marker of susceptibility to glioblastoma vs. other gliomas. On the other hand, the preliminary finding may be attained through chance alone since the observed association was not noteworthy (FPRP = 0.643) assuming a prior of 1%.In this current study, the genetic effect of -141GG genotype was observed for glioblastoma (WHO grade IV), but not for other gliomas. Our interpretation of the results is that different glioma subtypes or grades may be potentially involved with variable molecular and genetic events. Similar observations have been reported by Shete et al.  and SimoAPE1/Ref-1 has been implicated in the development and progression of various cancers [Ref-1 levels, is characteristic of adult gliomas [APE1/Ref-1 -141T/G variant was located in the promoter of the gene, only 141-bp upstream from the transcription initiation site. It is known that the promoter and 3'UTR regions of a gene have functions in regulating transcription, mRNA stability, and gene production [APE1/Ref-1 -141T/G polymorphism has been reported in lung cancer [APE/Ref-1 expression level or functions in glioblastom. Thus, more extensive studies are needed to resolve the functions of the -141T/G polymorphism. Nevertheless, we cannot at this time rule out the possibility that the genetic interactions between APE1/Ref-1 variants and other DNA repair genes are involved in glioblastoma. cancers -38. A production . Variantoduction ,41. The g cancer ,29. In tAPE1/Ref-1 polymorphisms and risk of gliomas or its subtypes only evaluated the nonsynonymous SNP rs1130409 [APE1/Ref-1 polymorphisms in glioma survival need to be further investigated.Previous studies on the association between s1130409 ,22; howes1130409  reporteds1130409 -47. Unfos1130409  reportedP value. Patients of GWA studies were whites from North America and European, and the allele and genotype frequencies described in the HapMap project for the rs1760944 polymorphism are different in CEU and CHB populations. However, these hypotheses can only be tested in GWA studies of gliomas in Chinese population.Two recent genome-wide association (GWA) studies with risk of gliomas did not identify an association between the rs1760944 polymorphism and glioma risk ,49. The AEP1/Ref-1 promoter variant was correlated with the AEP1/Ref-1 expression or its protein activity. Third, we only evaluated one SNP in the AEP1/Ref-1 gene, which was not sufficient to systematically evaluate glioma risk for the studied gene. It should be pointed out that our observed association could be due to linkage disequilibrium with the true causal SNP in or outside the AEP1/Ref-1 gene. In addition, we did not evaluate the gene-environment interactions in glioblastoma since the number of glioblastoma patients was small. However, replication of the finding should be warranted before proceeding to the gene-environment studies.Our study has several limitations. Mostly notably, the observed association in glioblastoma alone may be a subgroup finding due to chance. Therefore the finding requires validation in other independent cohorts. Secondly, our study did not provide any mechanistic information regarding glioma risk or protection. Specifically, we did not determine if the APE1/Ref-1 genotype in a Chinese Han population. Large-scale studies with ethnically diverse populations and functional evaluation are warranted to confirm our findings and to further elucidate the significance of the variant in glioblastoma.This study suggests a decreased risk of glioblastoma associated with a specific The authors declare that they have no competing interests.KZ collected the samples, performed the further interpretation of results, and drafted the manuscript. DH and JL carried out the data analysis generated the tables, as well as manuscript preparation. WF and CH participated in DNA sample preparation and co-worked in data analysis. GC also contributed to sample collection and face-to-face interview of each participant. RD and YM coordinated the study and participated in its design. QW and LZ and GD provided advices for preparing the manuscript and final editing of the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/11/104/prepub


MMP1 -1607 1G>2G polymorphism and risk of HNC, but their results have been inconsistent. Here, we conducted a meta-analysis to further explore the role of the MMP1 -1607 1G>2G polymorphism in HNC development.MMP1 is an important member of the MMP endopeptidase family that plays a critical role in the development of head and neck cancer (HNC). Several studies have investigated the association between the MMP1 -1607 1G>2G polymorphism and risk of HNC by calculating odds ratios (OR) and 95% confidence interval (CIs).We identified all eligible studies in the electronic databases of PubMed, ISI Web of Knowledge, MEDLINE, Embase, and Google Scholar (from January 2000 to June 2012). A meta-analysis was performed to evaluate the association between the Twelve studies were included in this meta-analysis. In overall comparison, significant associations were found using the recessive and allelic contrast models , but no association was detected using the dominant model. In the stratified analyses by several variables, significant associations were observed using the recessive, dominant, and allelic contrast models in the Asian population , European population , and population-based subgroup . Furthermore, significant associations were detected in oral cavity cancer and nasopharyngeal cancer under the recessive model.MMP1 -1607 1G>2G polymorphism is associated with risk of HNC and that it plays different roles in Asian and European populations. Further studies with large sample size are needed to validate our findings.Our results suggest that the  Head and neck cancer (HNC) globally comprises tumors of the oral cavity, nasopharynx, oropharynx, hypopharynx, and larynx, and is the sixth most common cancer in the world Matrix metalloproteinase-1 (MMP1) might serve as an important molecular marker for HNC. MMP is a family of zinc-dependent endopeptidases that are able to degrade essentially all extracelluar matrix (ECM) components, such as basement membranes, collagen, and fibronectin MMP1 gene is expressed in various kinds of normal cells, often at low levels under physiological conditions. However, MMP1 gene expression increases dramatically in a large number of malignancies, including HNC MMP1 (Collagenase-1), located on chromosome 11q22, is an important member of the MMP family that specifically degrades a major component of the ECM, type I collagen, as well as other fibrillar collagens of types II, III, V, and IX MMP1 plays a critical role in the regulation MMP1 gene transcription. Within this region, a functional single-nucleotide polymorphism (SNP), MMP1 -1607 1G>2G (rs1799750), has been identified MMP1 -1607 1G>2G contains a guanine insertion/deletion polymorphism at position -1607, which is relative to the transcriptional start site, and could result in higher expression of MMP1MMP1 -1607 1G>2G polymorphism and risk of HNC MMP1 -1607 1G>2G polymorphism in risk of HNC, we performed a meta-analysis by collecting and analyzing the genotyping data from all eligible case–control studies published to date.The promoter region of MMP1 polymorphism and risk of HNC (between January 2000 and June 2012), we conducted key word searches in the electronic databases of PubMed, ISI Web of Knowledge, MEDLINE, Embase, and Google Scholar. The key words we used included “head and neck cancer”, “oral cancer”, “pharyngeal cancer”, “hypopharyngeal cancer”, or “laryngeal cancer”, and “MMP1”, “Matrix Metalloproteinase1”, “collagenase”, and “polymorphism”, “variant”, “genotype”, or “SNP”. We also performed a manual search of the references of all identified articles in order to find additional studies. If important data were not reported in the original articles, we would contact with authors directly. Abstracts, unpublished reports and articles written in non-English languages were excluded.To identify all eligible case-control studies that examined the association between All data extraction was performed by two independent investigators, and they reached a consensus on all items by discussion. The following information was extracted from each included study: first author, published year, ethnicity of study population (Asian or European), numbers of case and controls, genotype distribution, genotyping methods, allele, etc.MMP1-1607 1G>2G polymorphism and risk of head and neck cancer, (2) case-control studies, (3) studies with sufficient available data to calculate an odds ratio (OR) with a 95% confidence interval (CI) and P -value, and (4) studies published in English. Studies with insufficient information about genotype frequency or number were not included. If the same population was included, with overlapping data, in more than one study, only the most recent or complete study was included in the meta-analysis.Eligible studies met the following criteria: (1) studies on the association between the MMP1-1607 1G>2G polymorphism and risk of HNC was assessed by OR and 95% CI. Heterogeneity between studies was evaluated by Cochran’s χ2 -based Q statistic test. If the P -value for heterogeneity was <0.05, suggesting that there was obvious heterogeneity of the data, a random-effects model was used; otherwise, we used a fixed-effects model to pool the results. The I2 test was also used to estimate the extent of heterogeneity between studies. As a guide, I2 values of <25% were considered “low”, value of ∼50% were considered “moderate”, and values of >75% were considered “high” MMP1-1607 1G>2G polymorphism and risk of HNC using a recessive genetic model (2G/2G versus 1G/1G+1G/2G), a dominant genetic model (2G/2G+1G/2G versus 1G/1G), and an allelic contrast model . In addition to an overall comparison, we also performed subgroup analyses based on the ethnicity of the study population, the source of the controls, and tumor site. The significance of pooled ORs was detected by the Z test (P<0.05 was considered significant). We investigated publication bias using the funnel plot and Egger’s test. Sensitivity analyses were also applied to assess the stability of the results by repeating the meta-analysis, omitting each study one at a time. All P values were two-sided, and all statistical analyses were carried out using STATA 11.0 software .First, we tested Hardy-Weinberg equilibrium (HWE) by comparing the expected and observed genotype frequencies of the control group using the Pearson chi-square test for goodness of fit. The association between the MMP1; 4 studies did not have sufficient data for further analysis; 4 studies were review articles; and 1 study was an article of comment. In a case-control study investigating the MMP1-1607 1G>2G polymorphism in two independent populations MMP1-1607 1G>2G polymorphism and risk of HNC   . Similarly, in an analysis of HWE studies, significant associations were found using the recessive model  and the allelic contrast model , and no significant association was observed using the dominant model .As shown in y<0.001)  and the y<0.001) , but no MMP1 -1607 1G>2G polymorphism had significant protective effects on the risk of HNC in all three genetic models , whereas in Asian population, it increased the risk of HNC significantly in all three models   , 3.MMP1 -1607 1G>2G polymorphism and risk of HNC in the population-based subgroup . However, no significant association was found in the hospital-based subgroup .Stratification based on the source of the controls showed significant associations between the 2, 70.9%, heterogeneityP = 0.001) and nasopharyngeal cancer .In the stratified analysis based on tumor site, significant associations were found in the recessive model for oral cavity cancer  or laryngeal cancer .However, no significant association was found for either pharyngeal (oropharynx/hypopharynx) cancer  and decreased significantly in the Asian population under the dominant model . Similarly, stratification based on the source of the controls significantly reduced the heterogeneity in the population-based subgroups .In this study, significant heterogeneity was found in all three genetic models. However, when the population were stratified by ethnicity, heterogeneity disappeared in the European population  and reported that the 2G allele increased significantly in OSCC patients when compared to controls, indicating that the MMP1 -1607 1G>2G polymorphism may be associated with risk of OSCC in a Chinese population MMP1 -1607 1G>2G and risk of OSCC in a Japanese population and found that the frequency of 2G alleles was significantly higher than that of 1G allele in OSCC patients MMP1 2G allele might play a crucial role in the early onset of OSCC. However, Zhou et al. reported that no significant association between the MMP1 -1607 1G>2G polymorphism and risk of HNC was found in two different populations A few of the source studies also reported results linking the 2G allele to an increased risk of HNC. O-charoenrat et al MMP1 -1607 1G>2G polymorphism played different roles in Asian and European populations. In the European population, it was significantly associated with reduced risk in all three genetic models in European population, whereas in Asian population, it was significantly associated with increased risk. For example, Zinzindohoue et al. examined the association of the MMP1 -1607 1G>2G polymorphism with risk of HNC in a case-control study in a European population MMP1 -1607 1G>2G polymorphism was associated with a decreased risk of oral cancer in 2G allele carriers in a European population MMP1 -1607 1G>2G polymorphism was associated with an increased risk of HNC in patients with the 2G/2G genotype or 2G allele carriers Interestingly, our subgroup analysis by ethnicity showed that the MMP1 genotype distributions in population-based controls may be similar to normal, and thus, population-based controls could be more reliable than hospital-based controls. This might partially explain why the results from the stratified analysis by the source of the controls were different between the two subgroups. In addition, another reason for the heterogeneity between studies was the tumor site. In the stratified analysis by tumor site, significant associations were found for oral cavity cancer and nasopharyngeal cancer, but not for either pharyngeal (oropharynx/hypopharynx) cancer or laryngeal cancer. Although HNC includes tumors from different sites, risk factors for these cancers are different. For example, oral cavity and laryngeal cancers are majorly associated with tobacco use and alcohol consumption, while oropharyngeal and nasopharyngeal cancers are principally related to viral infection, such as human papillomavirus (HPV) and Epstein-Barr virus (EBV). Thus, further studies with larger sample size and different tumor sites are warranted.Heterogeneity is a major problem when interpreting the results of meta-analyses. In this study, significant heterogeneity was detected in overall comparisons using all three genetic models. Ethnicity was an important reason for this heterogeneity. Individuals from different ethnicities may have diverse genetic backgrounds and environmental factors, and consequently, the same polymorphism may play different roles in different populations. Therefore, when we performed stratified analysis by ethnicity, the heterogeneity disappeared in the European population and decreased significantly in the Asian population. Furthermore, the source of the controls was another factor that contributed to heterogeneity. The The present study has some limitations. First, the study number was limited and the total sample size was relatively small; thus, our estimates of association might have occurred by chance. Second, significant heterogeneity was detected in our study, and thus, the results must be interpreted with caution. However, heterogeneity disappeared in some subgroups when stratified analysis was performed. Therefore, the results from the subgroup analyses may be more meaningful, as the polymorphism may play different roles in diverse subgroups. Third, further subgroup stratification based on other risk factors such as alcohol consumption, tobacco smoking and HPV status could not be performed because of the limited data MMP1 -1607 1G>2G polymorphism is associated with HNC risk. Moreover, subgroup analysis based on ethnicity indicates that it may play different roles in Asian and European populations. However, due to the limited study numbers and relatively small sample sizes, our results should be validated in future studies with larger sample sizes and in different ethnic populations.In conclusion, our meta-analysis suggests that the Checklist S1PRISMA 2009 Checklist.(DOC)Click here for additional data file.


Although the relationship between allergy and cancer has been investigated extensively, the role of allergy in head and neck cancer (HNC) appears less consistent. It is not clear whether allergies can independently influence the risk of HNC in the presence of known strong environmental risk factors, including consumption of alcohol, betel quid, and cigarette.The current paper reports results from: 1) an original hospital-based case-control study, which included 252 incident cases of HNC and 236 controls frequency-matched to cases on sex and age; and 2) a meta-analysis combining the results of the current case-control study and 13 previously published studies .In the original case-control study, we observed a strong inverse association between allergies and HNC . The meta-analysis also indicated a statistically significant inverse association between HNC and allergies , particularly strong for allergic rhinitis . In addition, the inverse association between allergies and HNC was observed only among men  but not among women .These findings suggest that immunity plays an influential role in the risk of HNC. Future studies investigating immune biomarkers, including cytokine profiles and genetic polymorphisms, are warranted to further delineate the relationship between allergies and HNC. Understanding the relationship between allergies and HNC may help devise effective strategies to reduce and treat HNC. Head and neck cancer (HNC), including cancers of the oral cavity, oropharynx and larynx, is one of the leading cancers worldwide. Each year, approximately 400,000 cases of oral and oropharyngeal cancer and 160,000 cases of laryngeal cancer are diagnosed worldwide The occurrence of most HNC can be attributed to the consumption of alcohol, cigarette, and betel quid Immune reactions manifested in the form of allergies have been extensively studied for their relationship with cancer risk Two hypotheses have been proposed to explain the inverse association between allergies and cancer, the “immunosurveillance hypothesis” and the “prophylaxis hypothesis” This hospital-based case-control study was approved by the institutional review boards of the National Cheng Kung University Hospital and the National Health Research Institutes. All participants of the study signed a written informed consent.All of the study subjects were recruited from the Department of Otolaryngology and the Department of Stomatology at the National Cheng Kung University Hospital from September 1, 2010 to June 30, 2012. The case subjects were patients newly diagnosed with HNC, including cancers of the oral cavity, oropharynx, and larynx, had no cancer history, aged 20 to 80 years, and had the ability to give informed consent. All HNC cases were pathologically confirmed squamous cell carcinoma. The controls subjects were patients who required surgery for non-cancerous diseases unrelated to cigarette smoking, alcohol drinking, or area nut/betel quid chewing, had no cancer history, had the ability to give informed consent, and frequency-matched to the cases on age (±5 years) and sex.In-person interview was conducted with each study subject by a trained interviewer using a standardized questionnaire. Demographic information, including sex, age, and educational level were collected. Detailed information (dose and duration) of the three major risk factors of HNC, including alcohol, betel quid, and cigarette, was ascertained. Study subjects were asked whether they had any of the five types of allergies, including allergic rhinitis, skin allergy, food allergy, drug allergy, and asthma. Those that responded positive were further asked about the age of onset for these allergies.To assess for the differences in demographic variables and lifestyle habits  between cases and controls, t-tests and chi-squared tests were performed for continuous variables and categorical variables, respectively. Multivariable unconditional logistic regression was performed to estimate the odds ratio (OR) and 95% confidence interval (CI) of HNC associated with allergies adjusted for age, sex, education, alcohol drinking (frequency), betel quid use (pack-years), and cigarette smoking (pack-years). Although, no pack-year definition has been established for betel-quid use, in order to adjust for the amount and the duration of betel quid use, a pack-year definition similar to the pack-year of cigarette of smoking was used, with 1 pack-year of betel quid use = 1 pack of betel quid (20 betel quids) use per day ×1 year. The analysis was first performed with all allergies combined . Subsequent analyses were performed to assess the association between HNC risk and individual types of allergies. To evaluate for the influence of allergy on the association between alcohol, betel quid, or cigarette and HNC, unconditional logistic regression was performed stratified by allergy status (yes/no). The heterogeneity of the association between alcohol, betel quid or cigarette and HNC risk by allergy status was assessed by the log-likelihood ratio test comparing the unconditional logistic regression model with the interaction term  to the model without the interaction term.A literature search was conducted until July 18, 2012 using the PubMed database and Thomson Reuters Web of Science with the search terms: “cancer AND  AND risk”. A total of 1,463 articles were identified using the search terms . ArticleThe data extracted from each article included authors, year of publication, types of study (cohort or case-control studies), number of study subjects, variables used for matching and for statistical adjustments, and relative risk  with corresponding 95% CIs. For studies that presented only frequency data, RRs and 95% CIs were calculated using the exact method. If a study only reported RRs separately by gender, specific types of allergy or HNC sites, a meta-analysis with a fixed effects model was performed to combine stratum-specific RRs into an overall RR of HNC.2 statistic was used to evaluate inconsistencies between studies 2 = 0–24.9%, 25–49.9%, and 50% or more represent low, moderate and high heterogeneity between studies, respectively. The summary RR was generated by a random effects model using the DerSimonian and Laird method if the heterogeneity was moderate to high (I2≥25%) 2<25%). The analysis was first performed to generate an overall meta-RR measuring the association between all allergies combined and HNC risk. Additional analyses were performed by study design (cohort vs. case-control), by sex, by HNC sites, and by specific types of allergies . Sensitivity analysis was performed using the “metainf” command by excluding one study at a time to examine whether any individual study had a dominant effect on the summary RR. Publication bias was examined by viewing the funnel plot and by Egger’s regression asymmetry test STATA software  was used to perform analyses of the current meta-analysis. Summary RRs (meta-RRs) was generated by the “metan” command P = 0.05) (P<0.0001)  and 236 controls were successfully recruited with a participation of 74.3% and 86.5% for the cases and controls, respectively. The HNC cases were slightly older than controls  . HNC cas = 0.05) . Compare<0.0001) . Cases h<0.0001) .Having any allergy was associated with a reduced risk of HNC  . This reThe most common type of allergy reported by the study subjects was allergic rhinitis (20% of cases and 41% of controls), which was associated with a reduced risk of HNC  . A consiThe increased risk of HNC associated with alcohol drinking, betel quid chewing or cigarette smoking appeared stronger among those without history of allergies, although the tests for heterogeneity were not statistically significant .2 = 76.5%) (2 = 79.0%) that was nearly identical to the result that combined all 14 studies, which indicated that the possibility of publication bias was minimal.Combining the data from 14 studies (9 cohort and 5 case-control studies), the meta-analysis showed a reduced risk of HNC associated with having allergies  . Sensiti= 76.5%)  showed t2 = 74.0%). The combined analysis of 5 case-control studies with 4,017 HNC cases and 10,928 controls showed a statistically significant inverse association between allergies and HNC .The combined analysis of 9 cohort studies with 727,569 subjects with allergies and 550 HNC outcomes showed a 9% reduction in HNC risk associated with allergies, although not statistically significant  but not among women . In the analysis with specific types of allergies, the strongest reduced risk of HNC was observed for allergic rhinitis  and a weaker reduction in HNC risk was observed for skin allergy  and asthma . A reduced risk associated with allergies was observed across all three sites of HNC   ; howeverA statistically significant inverse association between allergies and HNC was observed in our hospital-based case-control study. Combining the results of our study and those from the 13 published studies, the meta-analysis also indicated a statistically significant inverse association between HNC and allergies, particularly for allergic rhinitis and predominantly among men.et al. observed that the inverse association between allergies and cancer was more commonly found among tissues or organ systems that interface with the external environment  than among those that do not ; this led them to conclude that the “prophylaxis hypothesis” may be the more appropriate explanation for the site-specific inverse association between allergies and certain cancers The inverse association between allergies and cancer may be explained by two hypotheses, the “immunosurveillance hypothesis” and the “prophylaxis hypothesis” et al. examined the distribution of immunoglobulins in squamous cell carcinoma of head and neck by comparing tumor and normal tissues IgE plays a key role in most allergic reactions and evidence shows that IgE itself may possess anti-tumor activities et al. the association between allergies and HNC risk did not differ by smoking status  or drinking status  et al. In our original case-control study, there was a suggestion that allergies may attenuate the carcinogenic effect of alcohol, betel quid, or cigarette on the development of HNC risk, although the results were not statistically significant likely due to a sample size without sufficient statistical power. A possible explanation for the attenuating effect of allergies on HNC risk due to alcohol, betel quid, or cigarette may be a hyperactive immune function that can detect and destroy malignant cells before they have a chance to grow and expand. Alternatively, allergy-associated immune activities may rid body of the carcinogens before they have a chance to induce the oncogenic transformation of normal cells. However, in a previous study by Michaud e.g. common cold) as allergies. These recall errors were likely non-differential and would have biased the results toward the null; thus, the actual inverse association between allergies and HNC should be stronger in the absence of non-differential recall errors. Self-reported data in a case-control study can also be subject to differential recall bias. Usually, case subjects are more likely to ruminate about the possible causes of their diseases and may thus report more allergic symptoms, leading to a false positive association between allergies and HNC. However, the current meta-analysis indicated an inverse association between allergies and HNC for the case-control studies. In addition, the potential association between allergies and cancer is not well-known to the public; therefore, the inverse association between allergies and HNC among the case-control studies is unlikely due to differential recall bias. Compared to the cohort studies, the case-control studies were more likely to suffer from reverse causality, since the information on allergies was ascertained after the diagnosis of HNC and the immune function might be affected by the oncogenic process. In our original hospital-based case-control study, even when we considered allergies occurring more than 10 years before the diagnosis of cancer (or before the interview date for the controls), a strong inverse association between allergies and HNC still existed. In a hospital-based case-control by Vena et al. which considered only allergies diagnosed five years before the hospital admission, a reduced risk of HNC associated with allergies was observed among men e.g. cancer), is that there may not be enough outcomes during the follow-up period to produce a precise effect estimate The results from the meta-analysis showed that the inverse association between allergies and HNC was stronger among the case-control studies  compared to that among the cohort studies . All except one of the cohort studies on the association between allergies and HNC used data from medical records or clinical tests, whereas all of the case-control studies used self-reported data of allergies. Self-reported data might be more inaccurate due to recall errors associated with false memory or misinterpretation of non-allergic symptoms  can be attributed to alcohol and cigarette compared to women (50–61%) e.g. severity as indicated by the level of serum IgE, numbers of allergy-associated genetic variants, cytokine profiles) may provide a clearer picture to describe the relationship between allergies and HNC.Several limitations must be considered when interpreting the results of the current analysis. Because our case-control study is hospital-based, it is difficult to determine whether the cases and controls arise from the same source population. Our recruiting hospital is a major medical center in Southern Taiwan, and over 95% of our study subjects came from the county where the hospital is located. Even though we tried to exclude controls with diseases related to the use of alcohol, betel quid, and cigarette smoking, the possibility remains that our controls may not be representative of the background population for the cases. For the meta-analysis, studies were heterogeneous in design, patient population, disease sites, and types of allergies assessed. We explored the sources of heterogeneity by performing subgroup analyses by study design, disease sites, types of allergies, and gender. In addition, random-effect models were used to account for the heterogeneity across studies. Finally, the association between allergies and HNC was only assessed qualitatively with a binary status (yes vs. no) in the current analysis. More studies with quantitative measures of allergies  included in the meta-analysis for the association between allergies and head and neck cancer.(DOC)Click here for additional data file.Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.


The purpose of this meta-analysis was to evaluate the relationship between blood lutein and zeaxanthin concentration and the risk of age-related cataract (ARC). MEDLINE, EMBASE, ISI and Cochrane Library were searched to identify relevant studies up to April 2013. Meta-analysis was conducted to obtain pooled relative risks (RRs) for the highest-versus-lowest categories of blood lutein and zeaxanthin concentrations. One cohort study and seven cross-sectional studies were included in the meta-analysis. There were significant inverse associations between nuclear cataract and blood lutein and zeaxanthin concentrations, with the pooled RRs ranging from 0.63 : 0.49, 0.77) for zeaxanthin to 0.73  for lutein. A stronger association between nuclear cataract and blood zeaxanthin might be noted for the studies conducted in the European Nations. Blood lutein and zeaxanthin were also noted to lead towards a decrease in the risk of cortical cataract and subcapsular cataract; however, these pooled RRs were not statistically significant, with the exception of a marginal association between lutein and subcapsular cataract. Our results suggest that high blood lutein and zeaxanthin are significantly associated with a decrease in the risk of nuclear cataract. However, no significant associations were found for ARC in other regions of the lens. Cataract is a clouding or opacification of the lens inside the eye that obstructs the passage of light . Age-relLight-initiated oxidative damages are hypothesized to be the mechanism involved in ARC . The xanThis meta-analysis was performed in accordance with Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines .We searched MEDLINE, EMBASE, ISI Web of Science and Cochrane Library for relevant articles up to April 2013. Following MeSH words and text words were used: “lutein”, “zeaxanthin”, “xanthophyll”, or “carotenoid”, together with each of the following words “cataract”, “age-related cataract”, “ARC”, or “lens opacities”. Language restrictions were not imposed. Besides, manual searches of references cited by the retrieved articles were performed for additional literature. Authors and experts in the field were also contacted for relevant unpublished studies.All retrieved articles were examined by scanning their titles and abstracts and any clearly irrelevant studies were excluded. Full texts of the remaining articles were further reviewed. Studies were included in this meta-analysis if they met the following criteria: (1) the primary outcome was clearly defined as ARC; (2) the exposure of interest was blood (plasma or serum) levels of lutein and zeaxanthin; (3) odds ratio (OR) or relative risk (RR) estimates and their 95% confidence intervals (CIs) were provided; and (4) potential confounders were controlled for by matching or multivariable analysis in the studies. If a series of articles were published from the same study, the report with the most updated data was selected for analysis. Literature search and screening were conducted independently by three investigators  in a standardized manner, and any disagreements were resolved by discussion.Information from eligible studies was abstracted independently by three investigators , and discrepancies were resolved by consensus. We extracted the name of the first author, year of publication, country of origin, study design, characteristics of the study population , diagnostic method of ARC, classification and grading systems for ARC, blood concentration assessment methods, fully adjusted odds ratio or relative risk, and the adjustment factors. If a study provided several risk estimates, the most completely adjusted estimate was extracted.I² statistic were performed to evaluate statistical heterogeneity. Forest plots were used to visually assess the pooled ORs estimates across studies. Sensitivity analyses were performed to examine the influence of each individual study by iteratively excluding each study from the meta-analysis and comparing the point estimates including and excluding the study to determine their magnitude of influence on the overall estimate. Stratification and meta-regression analyses were performed to explore the source of heterogeneity among variables, such as country of origin, mean age of participants, method of ARC diagnosis and classification criteria of ARC. We assessed the potential of publication bias by using a funnel plot, Egger regression asymmetry test and Begg adjusted rank correlation test [For all studies, RRs or ORs and the corresponding 95% CIs were extracted from each article. Inverse variance-weighted risk estimates for ARC comparing the highest-versus-lowest categories of lutein and zeaxanthin were pooled with the fixed effects models or DerSimonian-Laird random effects models. The DerSimonian and Laird’s Q statistic and ion test ,17. All Our search resulted in a total of 1615 references, of which 1558 were excluded after abstract review . Of the I2: 0.0%; P: 0.54). For the individual xanthophyll carotenoids, the associations were consistently in the protective direction; however, blood zeaxanthin  tended to have a stronger association with decreasing the incidence of nuclear cataract than blood lutein . Exclusion of the cohort study essentially did not alter the results, with pooled RRs of 0.73  for lutein .The funnel plots for the studies evaluating blood lutein and zeaxanthin concentrations and their association with different subtypes of ARC revealed symmetry. The Egger and Begg tests suggested the absence of significant publication bias  than that in the plasma (0.2:1), suggesting that a selective uptake of zeaxanthin may occur in the lens tissue ,28. MoreThe stratified analyses showed that the association between nuclear cataract and blood zeaxanthin was significantly stronger for Europeans than it was for the Indians. This might be partly explained by the difference in zeaxanthin status for different populations. Compared with the Indians, blood zeaxanthin in Europeans was relatively higher ,21,22. MWith respect to cortical cataract and subcapsular cataract, we found no significant associations of these types of cataract with blood zeaxanthin or lutein. The difference in associations between cataract subtypes and serum zeaxanthin and lutein was probably due to the distinct pathogenesis for each type of ARC. With increasing age, a lower percentage of reduced glutathione can reach the lens nucleus, which makes the nucleus become less able to repair oxidative damage ,35. In cThe present study also has several potential limitations. First, our study is a meta-analysis of observational studies because of no randomized controlled trials (RCTs) found. It is commonly considered that observational studies are more likely to be subject to confounding and bias than RCTs, which can confound the results of our analysis. Second, the relatively small number of trials available evaluated the relationship between lutein and zeaxanthin status and the risk of cortical cataract or posterior subcapsular cataract; thus, our analysis was unlikely to detect substantially significant effect of xanthophyll carotenoids on opacities in these regions. Third, it may be that persons who have high blood levels of lutein and zeaxanthin are more likely to adhere to a healthier diet or lifestyle, which is associated with lower risks of ARC. Most studies adjusted for some major potential confounders; however, it was possible that residual or unknown confounding might bias the overall results in either direction. Fourth, although most subjects in studies from the European Nations were Europeans, it is possible that these study groups included participants of Asian and other non-European participants, which might lead to a mis-estimation of the association. Finally, although there was little evidence of publication bias from our funnel plot and rank correlation analysis, this possibility could be not completely excluded.Our meta-analysis demonstrated that increased blood concentrations of lutein and zeaxanthin might be associated with a reduced risk of nuclear cataract. However, there is insufficient evidence to support a significantly inverse relationship between blood lutein or zeaxanthin level and risk of other subtypes of ARC. Meanwhile, there are only a few observational studies published about the association between blood concentrations of these carotenoids and ARC, which also limits the power of meta-analysis. Therefore, longer-term, large-scale, prospective controlled intervention studies are needed to clarify further the impact of lutein and zeaxanthin on the development of ARC.


Elevated serum IL-6 level is a risk factor for coronary heart disease (CHD). The −174G>C and −572G>C polymorphisms in the IL-6 gene have previously been shown to modulate IL-6 levels. But the association between the −174G>C and −572G>C polymorphisms and the risk of CHD is still unclear. A meta-analysis of all eligible studies was carried out to clarify the role of IL-6 gene polymorphisms in CHD.PubMed, EMBASE, Vip, CNKI and CBM-disc were searched for eligible articles in English and Chinese that were published before October 2010. 27 studies involving 11580 patients with CHD and 17103 controls were included. A meta-analysis was performed for the included articles using the RevMan 5.0 and Stata 10.0 softwares. Overall, the −174C allele was not significantly associated with CHD risk  when compared with the −174G allele in the additive model, and meta-analysis under other genetic models  also did not reveal any significant association. On the contrary, the −572C allele was associated with a decreased risk of CHD when compared with the −572G allele . Furthermore, analyses under the recessive model  and the allele contrast model  yielded similar results. However, statistical significance was not found when the meta-analysis was restricted to studies focusing on European populations, studies with large sample size, and cohort studies by using subgroup analysis.The −174G>C polymorphism in the IL-6 gene is not significantly associated with increased risks of CHD. However, The −572G>C polymorphism may contribute to CHD development. Future investigations with better study design and large number of subjects are needed. Coronary heart disease (CHD), including myocardial infarction (MI), angina pectoris, and aterios sclerosis of the coronary arteries, is still one of the leading causes of death and disability. CHD also accounts for a huge consumption of health care resources especially in rapidly ageing societies The circulating concentrations of IL-6 are likely to be influenced by several environmental and genetic factors, subsequently affecting an individual's susceptibility to disease P = 0.05) in univariate analysis Several polymorphisms in the IL-6 promoter have been observed in healthy and unhealthy populations. Two single-nucleotide polymorphisms, the common −174 G>C variant In summary, previous researches investigating the relationship between the −174G>C or −572G>C polymorphisms and CHD risk in different study populations have yielded varied results. In order to clarify the role of IL-6 gene polymorphisms in CHD, we have conducted a formal meta-analysis for all eligible studies published before October 2010.Computer-based searches of PubMed, EMBASE, and China National Knowledge Infrastructure (CNKI), VIP database for Chinese Technical Periodicals (VIP), China Biological Medicine Database (CBM-disc) were performed without restricting language while using key words relating to CHD disease  in combination with words related to IL-6  and polymorphism . The searches included all articles published before October 2010. In addition, the reference lists in all relevant studies and review articles were also examined.Only studies that evaluated the association between CHD and the IL-6 gene polymorphisms were included. Included studies also met the following criteria: (1) They were original studies containing independent data, (2) the study design was case-control study or cohort study, (3) the diagnosis of disease was definite , (4) the −174G>C or −572G>C genotype frequencies of the IL6 gene were provided, (5) the risk of CHD was evaluated, and the odds ratio (OR) value could be calculated. The major reasons for exclusion were (1) they were family studies, (2) insufficient data about CHD, (3) lack of accessibility to original articles and duplicate publications.Two investigators independently reviewed all studies and extracted the data by using a standardized form. The following information was independently summarised from each included study: the first author, publication date, age, gender ratio, origin of country, source of controls, study design, ethnicity, definition of disease in cases, numbers of cases and controls, genotyping method, frequency and numbers for each genotype. Two reviewers compared the results from all included studies for accuracy and discussed any discrepancies before reaching to an agreement. If there were still some studies that the two reviewers failed to agree on, a third reviewer was then introduced to assess the studies.2-test for the control groups of each study. If the control groups were not in HWE, sensitivity analysis or subgroup analysis was performed to test the robustness of the findings.Deviation from Hard-Weinberg Equilibrium (HWE) was measured by χMeta-analysis was performed to investigate the association between the −174G>C and −572G>C polymorphisms in the IL-6 gene and the risk of CHD. We systematically assessed the risk of different genotypes of the −174G>C and −572G>C polymorphisms in CHD under various genetic models, which included the additive , the recessive (CC versus GG+GC), the dominant (GG+GC versus CC), the allele contrast (CC versus GG) and (GC versus GG) models. The strength of association between the −174G>C or −572G>C polymorphisms and CHD risk was measured by overall odds ratios (ORs) with 95% confidence interval (95%CI).I2PHeterogeneity≥0.1), the results were pooled using a fixed effect model and the Mantel-Haenszel method. Otherwise, a random-effect model and the Dersimonian and Laird method The homogeneity between studies was assessed using Q statistic test P-values were two-sided.All analyses were performed by using the Review-manager 5.0.1  and the Stata version 10.0  softwares. All A total of 486 articles were found based on the above searching criteria. 202 overlapping articles were excluded after comparing the title and author name. 220 articles were further excluded including reviews, editorials, and studies with insufficient detail on IL-6 or irrelevant outcome with CHD or MI by reading abstracts. Detailed evaluation on the full texts of the remaining 64 articles subsequently excluded 31 articles that failed to meet the inclusion criteria. Further assessment excluded 6 ineligible articles because of duplication In the 27 included articles, a total of 11580 patients and 17103 controls were investigated. The characteristics of those studies in the meta-analysis were listed in I2 = 35%, P = 0.036) was found. The overall OR under a random-effects model was 1.04 (95% CI = 0.98 to 1.10), and was not significant . Substantial heterogeneity among the studies  . Under tAccording to study characteristics, the subgroup analysis and sensitivity analysis were performed under the additive model. The results showed that the −572C allele in European population had no significant effect on the risk of CHD , whereas this effect was reversed in Asian population . Subgroup analysis by sample size suggested heterogeneous associations of the −572G>C polymorphism with CHD for the −572C allele in studies with large sample sizes , and a protective effect in studies with intermediate or smaller sample size . When stratifying by HWE, results did not change when restricted to the studies with the confirmed HWE. A sensitivity analysis comparing the fixed and the random models showed only minor changes in the estimates .2-residual = 28.26%, adj R2 = 62.12%) was a significant source of heterogeneity among studies (P<0.1). The other variables were not significant sources of heterogeneity, and all P values exceed 0.1  and ethnicity  were significant sources of heterogeneity among studies. The other variables were not significant sources of heterogeneity  and type of study. For the −174G>C genotype, univariate regression analysis showed that HWE status  value was 0.69 (0.51 to 0.93) for the −572G>C polymorphism after adjustment for ethnicity, and the association between the −572G>C polymorphism and CHD risk was not statistically significant after adjustment for other characteristics .t = 0.5, P = 0.619). P = 0.215).We used Egger's method to access the publication bias of the −174G>C and −572G>C polymorphisms in the IL-6 gene and CHD risk under the additive model. Atherosclerosis and CHD may be inflammatory conditions and IL-6 plays a key role in the inflammation process. A meta-analysis of prospective studies showed that the IL-6 plasma levels may be associated with CHD risk In the meta analysis of the −174G>C polymorphism, which involved 27812 subjects and 22 studies with 28 comparisons, significant association with CHD risk can not be found in overall comparisons under all genetic models. Results from studies with small sample size or deviating HWE are inconsistent. One explanation may be that small sample size and deviating HWE may cause results bias. Study characteristics, such as means age of cases, genotyping method, study design, source of controls and ethnicity, showed some differences in the included studies. But most of them were not responsible for heterogeneity by subgroup analysis, sensitivity analysis or meta-regression analysis.Another variant of the IL-6 gene, the −572G>C polymorphism, was recently reported Overall, a modest association existed between the 572G>C variant and CHD risk. But this association became non-significant when the meta-analysis was restricted to studies on European populations, studies with larger size and cohort studies. In addition, our meta-analysis does not support an association between the 174G>C polymorphism and CHD risk. Compared with a previous meta-analysis Several potential limitations of this meta-analysis should be considered. First, this meta-analysis only focused on papers published in English and Chinese. Second, not all the control subjects were age and sex matched to cases, which was likely one of the causes for heterogeneity. Third, the meta-regression analysis revealed that ethnicity and sample size were the significant sources of heterogeneity across studies on the 572G>C polymorphism and CHD risk. Moreover, because the included studies are observational, confounding and biases might have affected the pooled estimates. However, the advantages of this meta-analysis were also obvious. This study was based on a large sample size of 27 included studies with 11580 cases and 17103 controls. This is twice the sample size that had been reported in the previous meta-analysis. Second, the potential sources of heterogeneity in the meta-analysis were assessed in detail. Third, the associations of two variants of the IL-6 gene with risk of CHD evaluated under different genetic models were similar.In conclusion, the meta-analysis provides evidence that the −174G>C polymorphism in the IL-6 gene is not significantly associated with increased risk of CHD. A significant association can be found between the −572G>C polymorphism in the IL-6 gene and CHD risk, especially in Asian populations. Furthermore, well-designed large population studies are needed to investigate the relation of these polymorphisms and CHD.Figure S1Association between IL-6 gene −174G>C polymorphism and CHD risk under the additive genetic model (tiff).(TIF)Click here for additional data file.Figure S2The cumulate meta-analysis between IL-6 gene −174G>C polymorphism and CHD risk under the additive genetic model according to publication year (tiff).(TIF)Click here for additional data file.Figure S3Association between IL-6 gene −572G>C polymorphism and CHD risk under the additive genetic model (tiff).(TIF)Click here for additional data file.Table S1Basic characteristics of the included studies in the meta-analysis (DOC).(DOC)Click here for additional data file.Table S2Detailed characteristics of the included studies in the meta-analysis (DOC).(DOC)Click here for additional data file.Table S3The distribution of IL-6 gene −174G>C genotypes and alleles among cases and controls, and P-value of HWE in control (DOC).(DOC)Click here for additional data file.Table S4The distribution of IL-6 gene −572G>C genotypes and alleles among cases and control, and P-value of HWE in controls (DOC).(DOC)Click here for additional data file.Table S5The meta-regression analysis for heterogeneity under the additive model of IL-6 gene −174G>C polymorphism (DOC).(DOC)Click here for additional data file.Table S6The meta-regression analysis for heterogeneity under the additive model of IL-6 gene −572G>C polymorphism (DOC).(DOC)Click here for additional data file.Table S7ORs and 95%CI with or without adjusted factors for coronary heart disease and the −174G>C, −572G>C polymorphism in IL-6 gene under additive model (DOC).(DOC)Click here for additional data file.


Significant results were also observed for heterozygous  and homozygote . There was strong evidence of heterogeneity, which largely disappeared after stratification by ethnicity. After stratiﬁed by ethnicity, significant associations were found among Caucasians. However, no significant associations were detected among East Asian and African populations. In addition, we found that rs11249433 polymorphism on 1p11 confer risk, exclusively for ER-positive tumors with per-allele OR of 1.13  compared to ER-negative tumors of 1.01 . Similar results were also observed when stratified by PR status. Our findings demonstrated that rs11249433-G allele is a risk-conferring factor for the development of breast cancer, especially in Caucasians.Genome-wide association studies have identified SNP rs11249433 at chromosome 1p11 as a new breast cancer (BC) susceptibility locus in populations of European descent. Since then, the relationship between 1p11- rs11249433 and breast cancer has been reported in various ethnic groups; however, these studies have yielded inconsistent results. To investigate this inconsistency, we performed a meta-analysis of 15 studies involving a total of 90,154 cases and 137,238 controls for 1p11-rs11249433 polymorphism to evaluate its effect on genetic susceptibility for breast cancer. An overall random effects odds ratio of 1.09 (95% CI: 1.06-1.12, P<10 Breast cancer (BC) is one of the most common malignancies among women worldwide . EpidemiEligible literatures published before the end of March 2013 were identified by a search of PUBMED, EMBASE, and ISI web of science and CNKI  databases using combinations of the following keywords: “breast cancer,’’ ‘‘malignant breast neoplasm,’’ ‘‘1p11,’’ ‘‘rs11249433,’’ ‘‘polymorphism,’’ “variation”, without restriction on language. All references cited by identiﬁed eligible studies and previous reviews were scrutinized to ﬁnd additional work not indexed by PubMed.Articles were included in this meta-analysis if they (1) examined the hypothesis that 1p11-rs11249433 polymorphism was associated with breast cancer risk, (2) followed a case-control or cohort study design, (3) identified breast cancer cases histologically or pathologically, and (4) provided sufficient information on genotype/allele counts between cases and controls to estimate the odds ratio (OR) and the corresponding 95% confidence interval (95% CI).Information was carefully extracted from all eligible publications independently by two of the authors according to the inclusion criteria listed above. The following variables were extracted from each study: the ﬁrst author, published year, study design, geographic area, ethnicity, mean age of cases and controls, case-control match status, deﬁnition and numbers of cases and controls, source of controls, genotyping method, frequency of genotypes, and Hardy-Weinberg equilibrium (HWE) in controls. Relevant clinical characteristics included estrogen receptor (ER) status, progesterone receptor (PR) status, and tumor grade. Review reports from the two were then compared to identify any inconsistency, and differences were resolved by further discussion among all authors. Studies with different ethnic groups were considered as individual studies for our analyses.For association studies with inconsistent results on the same polymorphisms, the methodological quality should be assessed by appropriate criteria to limit the risk of introducing bias into meta-analyses or systematic reviews. A procedure known as ‘extended-quality score’ has been developed to assess the quality of association studies. The procedure scores each paper categorizing it as having ‘high’, ‘median’ or ‘poor’ quality. Detailed procedure of the quality assessment was previously described .2 tests with 1 degree of freedom. The strength of association between 1p11-rs11249433 polymorphism and breast cancer risk was assessed by OR with the corresponding 95% CI. The per-allele OR of the risk allele was compared between cases and controls. Then, we estimated the risks of the heterozygous and homozygous genotypes on BC compared with the wild-type homozygote. Cochran’s Q statistical test and I2 were performed to assess possible heterogeneity between the individual studies, and thus to insure that each group of studies was suitable for meta-analysis . However, no significant association was detected for ER- tumors   = 0.99], compared to PR- tumors with per-allele OR of 1.04  . SimilarSensitivity analysis was performed by excluding one study at a time. The results confirmed the significant association between the rs11249433 polymorphism and the risk of BC, with ORs and 95% CIs ranging from 1.08 (95% CI: 1.05-1.12) to 1.10 (95% CI: 1.07-1.13). Funnel plot and Egger’s test were performed to evaluate the publication bias of the literature reviewed. The shape of the funnel plots seemed symmetrical, suggesting no publication bias among the studies included . The EggThe pathogenesis of the development and progression of breast cancer is far from being clear at present. Accumulated evidence suggests that it is a complex polygenic disorder for which genetic factors play an important role in disease etiology . This isIn the subgroup analysis, ethnicity was responsible for heterogeneity, and the ORs between different genetic models and sample size were consistent. The rs11249433 showed a positive association with BC in Caucasians; whereas no associations were found in East Asians and populations of African descent. There are several possible reasons for such ethnic differences. Firstly, the G allele frequency among controls was 0.03 in East Asians, 0.12 in Africans and 0.40 in Caucasian population, suggesting a possible role of ethnic differences in genetic backgrounds. Therefore, failing to identify any significant association in East Asians and Africans could be due to substantially lower statistical power caused by the relatively lower prevalence of G allele of 1p11-rs11249433. Secondly, study design or small sample size or some environmental factors may affect the results. ER status may be particularly important given that some GWAS findings are specific to ER-positive and ER-negative cancers ,35 and bMeta-analysis is often dominated by a few large studies, which markedly reduces the evidence from smaller studies. By considering sample size, significantly increased BC susceptibility in 1p11 rs11249433 variation was also found both in large and small studies for all genetic models. However, our results suggest an overestimation of the true genetic association by small studies.ER status is known to affect prognosis of BC. Stratiﬁcation of tumors by ER status indicated that the 1p11 rs11249433 confer risk, preferentially for estrogen ER-positive tumors, with no risk for ER-negative BC. Results from subgroup analyses on ER status of tumors were in agreement with previous reports ,38. ThisThe mechanism underlying the association of the 1p11 rs11249433 polymorphism with BC risk remains unknown. Recently, a study conducted by Fu et al. demonstrated that the expression of NOTCH2 differs in subgroups of breast tumors and by genotypes of the breast cancer-associated SNP rs11249433 . The NOTSeveral potential limitations of the present meta-analysis should be taken into consideration. Firstly, our results were based on unadjusted estimates, while a more precise analysis should be conducted if all individual-level raw data were available, which would allow for the adjustment by other co-variants including age, cigarette consumption, alcohol drinking, menopausal status, and other lifestyle. Secondly, the subgroup meta-analyses on East Asian and African populations are based on a small number of studies with such information available. Nevertheless, the total number of subjects included in this part of the analysis comprises the largest sample size so far. As studies among the Non-Caucasians are currently limited, further studies including a wider spectrum of subjects should be carried to investigate the role of these variants in different populations. Thirdly, the single locus–based nature of meta-analysis precluded the possibility of gene-gene and gene-environment interactions, as well as haplotype-based effects, suggesting that additional studies assessing these aspects are necessary.To the best of our knowledge, this study was the ﬁrst comprehensive meta-analysis to assess the relationship between the 1p11-rs11249433 polymorphism and BC susceptibility. Our meta-analysis showed that rs11249433 polymorphisms at 1p11 might be risk-conferring factor for the development of BC in Caucasians, but not in East Asians and Africans. As studies among these populations are currently limited, further studies including a wider spectrum of subjects to investigate the role of this variant in these populations will be needed.Figure S1The flow chart of the included studies.(TIF)Click here for additional data file.Figure S2Per-allele odds ratios and 95% conﬁdence intervals for the association between 1p11-rs11249433 and BC risk by PR status.(TIF)Click here for additional data file.Figure S3Funnel plot of 1p11-rs11249433 polymorphism and BC risk.(TIF)Click here for additional data file.Checklist S1(DOC)Click here for additional data file.


However, moderate levels of heterogeneity were found between study-level effect sizes. Meta-regression analyses indicated that studies with older participants that used SRT tasks with second order conditional sequences, or with older participants that used sequences that were presented a large number of times, were associated with smaller effect sizes. These associations are discussed with respect to compensatory and delayed memory systems in dyslexia.A number of studies have investigated procedural learning in dyslexia using serial reaction time (SRT) tasks. Overall, the results have been mixed, with evidence of both impaired and intact learning reported. We undertook a systematic search of studies that examined procedural learning using SRT tasks, and synthesized the data using meta-analysis. A total of 14 studies were identified, representing data from 314 individuals with dyslexia and 317 typically developing control participants. The results indicate that, on average, individuals with dyslexia have worse procedural learning abilities than controls, as indexed by sequence learning on the SRT task. The average weighted standardized mean difference (the effect size) was found to be 0.449 (CI DyslexiBehavioral investigations have revealed a pattern of deficits in dyslexia beyond reading impairments. Evidence suggests that difficulties with phonological processing may constitute the core impairment in dyslexia, in particular problems with phonological awareness, that is, the ability to identify and manipulate the sound structure of words in a language . Researc1.1Similarly, there has been ongoing interest in whether one or more functions of the procedural memory system also contribute or underlie the reading impairments in dyslexia . This meIt has been hypothesized that the reading impairments in dyslexia may be at least partly explained by problems with the procedural memory system. A key prediction of the proposal that procedural memory impairments are found in and may underlie dyslexia, is that individuals with dyslexia should in fact have worse procedural learning abilities than control individuals with typical reading skills. A number of studies have examined learning in procedural memory in dyslexia, using a variety of paradigms, including artificial grammar learning , alternaQualitative reviews cannot easily synthesize this literature, while accounting for study-specific features such as effect size, sample size, and task related methodological differences. Rather, a rigorous quantitative approach using meta-analytic techniques is more appropriate. In meta-analysis, the results from similar individual studies with similar methodologies are pooled, permitting population parameters to be estimated with greater precision .1.2This report used meta-analysis and meta-regression to evaluate and synthesize existing evidence and determine whether procedural learning is affected in dyslexia and what factors may influence the observed inconsistencies in the literature. We focused on the SRT task because this task has been the most widely used to examine procedural learning in dyslexia, with fully 14 studies to our knowledge, and indeed, is the best-studied procedural learning task more generally  control individuals. The key comparison in these studies is whether the difference in RTs between sequence and random blocks is significantly larger in TD individuals than in participants with dyslexia. Although a number of studies have indeed reported such findings , others A number of explanations could account for the inconsistent pattern of findings. One possibility, of course, is that procedural learning impairments are in fact not present in dyslexia, contrary to the predictions of Second, procedural learning impairments might indeed be reliably present across individuals in the disorders, and the heterogeneity of results is due to noise or to insufficient power  in some studies. On this perspective, a meta-analysis of SRT studies should reveal a deficit on this task, but no other variables  should account for any of the variability of findings across studies.Third, procedural impairments may occur primarily in certain dyslexic subgroups, and/or under certain testing conditions. For example, it has previously been suggested that inconsistent findings pertaining to other cognitive and motor impairments in dyslexia might be due to deficits restricted largely to certain subgroups e.g., . AdditioResearch indicates declarative memory improves throughout childhood . In concVariability in methodological conditions could also help explain learning variability in the SRT task in dyslexia. Multiple variants of this task have been employed in studies of dyslexia. The task varies in a number of respects. For instance, the length of training, for example, as measured by the number of exposures to the sequence (number of blocks times number of sequences per block), has varied substantially between studies, from as little as 10  to as muAlso variable between studies is the length of the sequence used in the SRT task. In the dyslexia literature, this ranges from as few as five items  to as matype of sequence. First order conditional (FOC) sequences and second order conditional (SOC) sequences have been used in past research into dyslexia  sequences involves the medial temporal lobes in addition to basal ganglia and cerebellum . In accoTo understand whether SRT task performance is indeed at least partly explained by the above factors, they need to be systematically examined. In this report we first used meta-analysis to systematically synthesize the evidence related to performance of individuals with dyslexia on SRT tasks. The purpose of this analysis was to investigate whether individuals with dyslexia have poorer procedural learning compared to typically developing control participants who did not have dyslexia. We then used meta-regression to investigate whether there were systematic influences that might account for inconsistent findings in the literature examining SRT in dyslexia. Specifically, we investigated whether participants’ age and characteristics of the SRT task  could account for discrepancies between studies.22.1We followed the methods used by previous meta-analyses of SRT task performance in other clinical groups, such as individuals with Parkinson's disease . Article2.2The inclusionary criteria used in this meta-analysis were based on the protocols used by 2.3After the removal of duplicate entries, one reviewer assessed all the abstracts. A random sample of 10% of all abstracts was assessed by a second reviewer. Any disagreements were resolved by discussion. Finally, the reviewers independently retrieved and screened full-text articles accordingly to the eligibility criteria. Inter-rater reliability was strong. For 42 out of the 43 articles (Cohen's kappa = .988) the reviewers independently agreed upon the suitability of each article for inclusion in the meta-analysis. Agreement about the suitability of one article was reached through consensus. A total of 14 published studies were included, and their data was extracted for the meta-analysis. A summary of each of study's participants and SRT task structure are summarized in 2.4The standard method for comparing the performance of two groups on an SRT task tests whether the difference in RTs between the final random block and preceding block comprising sequenced stimulus presentations differs between the study and control group e.g., . From eaFollowing pooled is within-group standard deviation of the difference between the final random block and preceding block, pooled across the control and study group.SDThe result from each study included in the meta-analysis was described using a single effect size. For 11 studies, a single effect size was extracted for each study . For thr2.5To address the first question, that is, whether there was a difference between individuals with dyslexia and TD controls on sequence learning in SRT tasks, effect sizes were pooled and a weighted averaged effect size was computed using a random effects model . A signiI2 statistic (I2 describes the amount of heterogeneity in effect sizes (as a percentage or proportion) that is attributable to between-study effects . Alternatively stated, the I2 statistic measures variability in effect sizes not attributable to sampling error. As a guideline The second question addressed in this report was whether different methodological characteristics accounted for differences in study level effect sizes. In undertaking this analysis we first measured the total amount of heterogeneity using the tatistic . I2 descFinally, meta-regression  was used33.1t (12) = 1.32, p = .212). This suggests the distribution of effect sizes is adequately symmetrical and therefore that publication bias is unlikely.Preliminary analyses investigated the presence of publication bias using a funnel plot, which plots a measure of study precision (using standard error) against individual study effect sizes . These d3.2p < .001). This indicates that on average, individuals with dyslexia perform around half a standard deviation worse than controls of a comparable age on sequence learning in SRT tasks. According to The first research question addressed whether or not individuals with dyslexia are worse at procedural learning than TD control individuals. The effect sizes computed for each study and the weighted average effect size are presented in I2 statistic indicated that 53.1% of variability between effect sizes represents true heterogeneity . According to the guidelines by Despite the finding that the overall effect size is statistically significant, inspection of 3.3Random-effects model meta-regression was used to investigate the second research question: whether the mean of age of participants in each study and methodological factors account for differences in study findings. The methodological factors examined were sequence type, sequence length and number of exposures to the sequence see . In the To have sufficient statistical power in meta-regression an effect size to covariate ratio of 10:1 is suggested see . Since tThe outcome variables used in the following meta-regressions were the effect sizes reported in Models 1–4 tested the contribution of age and methodological characteristics of the SRT task as main effects in accounting for differences in effect sizes; none were found to fit the data. However, two out of the six models testing an interaction were found to account for significant amounts of heterogeneity. Model 5 which tested the Age × Number of Exposures to the Sequence interaction term was found to be a significant predictor of effect sizes. This model accounted for 37% of between-study heterogeneity. This result indicates effect sizes become smaller in studies where participants are exposed to the sequence more times, but only when participants are older in age. In other words, the difference between the dyslexic and control groups on the SRT task decreases when more training to the sequence is provided and when participants are older. This model plotted against observed effect sizes is presented in Panel A in Model 6, which tested the Age × Sequence Type interaction term, was also found to be a significant predictor of effect sizes. This model accounted for 28% of between-study variance. This indicates that effect sizes become smaller for studies that presented SOC sequence, but only when participants were older. That is, the difference between dyslexic and control groups on the SRT task is smaller for SOC sequence than for FOC sequences, but only for older participants. Panel B in 4This report evaluated and synthesized available evidence of procedural learning deficits in dyslexia as indexed by sequence learning performance on SRT tasks. Following a systematic search of the literature, 14 studies were identified that were included in the meta-analysis. The weighted average effect size computed from these studies was found to be .449 (a medium effect size), which was statistically significant. This result indicates that on average, the mean difference in RTs between the final random block and preceding sequence block in SRT tasks is about half a standard deviation smaller in individuals with dyslexia than in typically developing control participants. The findings of this meta-analysis provide strong evidence in support of the hypothesis that procedural memory is impaired in dyslexia, and that this may help account for the reading deficits in the disorder .I2 statistic). As a reminder, in meta-analysis when using a random effects model to pool effect sizes, heterogeneity between effect sizes is assumed to reflect within-study and between-study variability. Within-study variability describes variability due to sampling error. Between-study variability refers to systematic influences on study effect size including participant or methodological characteristics. The observed I2 value of 53% indicates that 47% of heterogeneity between effect sizes reflected sampling error and 53% between-study error. Thus just over half of the observed heterogeneity in effect sizes in SRT studies in dyslexia appears to reflect systematic influences.Using meta-regression, we also investigated potentially influencing factors that could account for the inconsistency of the findings in the SRT literature in dyslexia. We observed moderate to high levels of heterogeneity (which was formally quantified using the Two meta-regression models accounted for significant between-study heterogeneity. In one model, an interaction term comprising Age by Sequence Type was found to be a significant predictor of study level effect sizes (Model 6; see An intriguing possibility we would like to emphasize relates to the significant ‘Age by Sequence Type’ interaction found. We suggest that this interaction might reflect compensatory mechanisms of the declarative memory system . As discThe interpretation of the meta-regression analysis showing a statistically significant ‘Age’ by ‘Number of Exposures to the Sequence’ interaction see  Model 5 4.1Two limitations need to be taken into account when interpreting the results presented in this report. First, results from the meta-analysis showing poorer procedural learning in dyslexia see  cannot a4.2In the ongoing effort to improve detection rates and remediation of reading difficulties, the results from the current study suggest procedural memory is worthy of further consideration. With respect to detection, results presented in this report indicate poor procedural memory might be a risk factor for reading problems. Further research is needed to develop a practical, time-efficient assessment of procedural memory that can be used in clinical and other remediation contexts. The extent to which focusing on procedural memory in the context of remediation can lead to improved reading outcomes is crucially dependent on whether a causal link exists between poor procedural learning and reading problems; a topic for future research. The findings of these meta-analyses also raise issues regarding the focus for remediation-related research. Should remediation harness the compensatory potential of declarative memory, should intervention be directed to procedural memory, or both? More generally, the interactions observed between age and task performance on study findings, serve as a timely reminder that neurodevelopment needs to be taken into account to better understand the underlying nature of dyslexia see .5In this report meta-analysis was used to evaluate the evidence for procedural learning deficits in individuals with dyslexia. The studies included in the meta-analyses measured procedural learning using SRT tasks. The weighted average effect size computed from 14 studies, representing data from 314 individuals with dyslexia and 317 control individuals, showed a significant difference between the groups in the SRT measure of procedural learning. We found that an interaction of age and methodological characteristics of the SRT task best accounted for differences between individual study findings. In sum, this report provides strong evidence of a procedural learning deficit in dyslexia. These reports also underline important areas for future research. In order to determine the clinical significance of impaired procedural memory studies are needed to examine potential causal links between this aspect of memory and reading difficulties.