Background. This meta-analysis investigated the association between functional COX-2 gene polymorphisms and the risk of oral cancer. Methods. Several electronic databases were searched for published studies using combinations of keywords related to COX-2 gene polymorphisms and oral cancer. After selection of relevant studies, following strict inclusion and exclusion criteria, data was performed using STATA 12.0 software. Results. We retrieved 83 studies from database search using specific search terms. After multiple rounds of selection and elimination, 7 studies were finally identified as suitable to be included in our present meta-analysis, based on their relevance and data integrity. These 7 studies contained a combined total of 2,296 oral cancer patients and 3,647 healthy controls. Our findings demonstrated that +837 T > C (rs5275) polymorphism in COX-2 showed statistically significant differences in gene frequencies in case and control groups in allele model and dominant model. Similar results were obtained with COX-2 gene polymorphism 765 G > C (rs20417). On the other hand, 1195 A > G (rs689466) polymorphism in COX-2 did not confer susceptibility to oral cancers. Conclusion. Based on our results, COX-2 gene polymorphisms, +837 T > C (rs5275) and −765G > C (rs20417), showed clear links with oral cancer susceptibility, and the 1195A > G (rs689466) polymorphism did not show such a correlation. Individ COX-2 is observed in many cancers, especially in the upper aerodigestive tract cancers, such as oral cancer, gastric cancer, and esophageal cancer, and is associated with cell proliferation, inhibition of apoptosis, tumor invasion, and angiogenesis [ COX-2 gene is located on chromosome 1 locus of 1q25.2–q25.3 and is 8.0 kbp in size with 10 exons [ COX-2 gene polymorphisms affect the expression levels and enzymatic activity of COX-2 and therefore are intimately linked to inflammatory response and individual variations in the susceptibility to oral cancers [ COX-2 gene, −1195 G > A , +837 T > C (rs5275), and −765 G > C (rs20417) have received considerable attention for their close links to oral cancers, compared to the other SNPs of COX-2 gene [ COX-2 expression, and the −765 G > C (rs20417) polymorphism is located in the 3′UTR and influences COX-2 mRNA stability and translation. Both polymorphisms are independently associated with several cancers and evidence shows that polymorphisms of COX-2 gene, in general, enhance cancer risk [ COX-2 genetic polymorphisms in cancers, especially in different ethnic groups [ COX-2 gene polymorphisms with the risk of oral cancer.COX-2, also named prostaglandin-endoperoxide synthase (PTGS), is a key enzyme in the arachidonic acid pathway, initiating the synthesis of biologically important prostanoids and eicosanoids . Overexpogenesis . The hum10 exons . COX-2 g cancers –15. ThreX-2 gene . The +83cer risk . However COX-2 gene polymorphisms and susceptibility to oral cancers, we exhaustively searched PubMed, EBSCO, Ovid, Springerlink, Wiley, Web of Science, VIP, Wanfang, and China National Knowledge Infrastructure (CNKI) databases , utilizing selected keywords related to oral Cancer, COX-2, and polymorphism genetic. The search terms applied in our literature search were as follows: “Mouth Neoplasms” or “Oral Cancer” or “Tongue Neoplasms” or “Gingival Neoplasms” or “Lip Neoplasms” or “Palatal Neoplasms” or “Salivary Gland Neoplasms” and “Cyclooxygenase 2” or “COX-2” and “Polymorphism, Genetic”. We also manually examined the bibliographies of relevant articles to identify additional studies.To identify all relevant studies on COX-2 gene polymorphism and oral cancer susceptibility; (3) the outcome index involved the allele gene and frequencies of genotype in both case and control groups; (4) when the same author published articles using the same clinical data, study with the largest sample size or the newest publication was used; (5) only studies published in English and Chinese are included. The major exclusion criteria were as follows: (1) containing summary and abstracts only; (2) nonhuman studies; (3) duplicate publications or unpublished studies; (4) no sufficient information provided.The following criteria were applied for literature selection to be included in the present meta-analysis: (1) study types should be case-control studies; (2) the study topic should be related tohttp://www.casp-uk.net/). The CASP criteria are scored as follows: the focused issue is clearly addressed (CASP01); the research problem is eligible and the research design answers the research problem (CASP02); the cases were enrolled acceptably (CASP03); the controls were selected acceptably (CASP04); the measurement for exposure factors is precise to minimize bias (CASP05); the study controls other crucial confounding factors (CASP06); the research results are complete (CASP07); the research results are precise (CASP08); the research results are reliable (CASP09); the research results are applicable to the local population (CASP10); the research results fit with other available evidence (CASP11).The data were extracted from each included study by two independent investigators, and the following information was collected: surname and initials of the first author, year of submission, country, ethnicity, language, age, gender, sample size, genotyping methods, study design, Hardy-Weinberg equilibrium test (HWE), SNP, and gene. Disagreement on the inclusion of any study was resolved by consultation with a third investigator. The quality of included trials was assessed utilizing the critical appraisal skills program (CASP) for case-control Studies  and I2 tests to quantify heterogeneity among studies [P < 0.05 or I2 > 50% indicated heterogeneity random effects model was used; otherwise fixed effects model was employed. In order to reflect the influence of single studies on the results the sensitivity analysis was employed. In addition, potential publication bias was examined by using funnel plots as well as Egger's linear regression test to ensure the reliability of results (P < 0.05 was considered significant) [All statistical tests for this meta-analysis were performed with STATA 12.0 . The relative risk (RR) and 95% confidence intervals (CI) were estimated by the fixed effects model or random effects model to evaluate the correlation betweenect size . We used studies . When P ificant) , 20.P = 0.006), other genotype distributions were in accordance with HWE (P > 0.05). CASP scores and baseline characteristics for eligible studies are presented in A total of 83 articles were retrieved through electronic database and manual searches. After rejecting 31 duplicate studies, 3 letters and reviews, 5 nonhuman studies, and 22 unrelated studies, the remaining studies were reviewed in full text for data integrity. This resulted in further elimination of 11 articles, along with 4 studies that lacked sufficient data . Eventua COX-2 gene polymorphism and the susceptibility to oral cancer. According to the heterogeneity test, the studies showed significant heterogeneity . Our findings demonstrated that +837 T > C (rs5275) polymorphism in COX-2 elevates the susceptibility to oral cancer and there were significant statistical differences in gene frequencies between the case and control groups in both allele model and dominant model  (Four studies reported the association between +837 T > C (rs5275)< 0.001) , Table 2 COX-2 related to the susceptibility to oral cancer. The result of the heterogeneity test indicated significant heterogeneity among various studies . Our analysis suggested that the COX-2 gene polymorphism 765 G > C (rs20417) is correlated with the susceptibility of oral cancers. The gene frequencies of case group and control group, under allele model and dominant model, exhibited significant differences   polymorphism in< 0.001) , Table 2 COX-2 linked with susceptibility to oral cancer. No significant heterogeneity was detected; thus fixed effects model was adopted . Our findings of meta-analysis showed that 1195 A > G (rs689466) polymorphism in COX-2 did not confer susceptibility to oral cancer. The gene frequencies of case group and control group under allele model and dominant model showed no significant differences  (Three studies reported that the 1195 A > G (rs689466) polymorphism in= 0.186) , Table 3P > 0.05), indicating highly reliable results  related to −765 G > C (rs20417) and Chiang (2008) linked to 1195 A > G (rs689466); the remaining studies had no influence on the estimated pooled RR . The res results .COX-2 is an inducible enzyme that catalyzes the conversion of arachidonic acid to prostaglandins, and the reaction products influence cell proliferation and are key mediators of inflammation [ COX-2 plays a key role in carcinogenesis by inhibiting apoptosis, promoting tumor growth, angiogenesis, invasion, and metastasis [ COX-2 in the etiology of oral cancers, genetic variations of COX-2 gene affect the susceptibility to cancer development [ COX-2 +837 T > C (rs5275) and −765 G > C (rs20417) variant alleles were associated with significantly increased risk of oral cancer, while the effects of 1195 A > G (rs689466) need further exploration. The guanine (G) to cytosine (C) conversion at position −765 bp lies in the promoter region of COX-2 gene, and the −765 G > C polymorphism affects transcription activity and is the most extensively studied COX-2 polymorphism [ COX-2 mRNA is influenced by 3′UTR elements, and the exon 10 +837 T > C SNP is located in the 3′UTR and alters mRNA stability and translation efficiency to influence susceptibility to oral cancer [ammation . Evidenctastasis , 27–29. elopment . In our morphism . COX-2 pmorphism . Furtherl cancer . Thus, tl cancer . Our conl cancer .Limitations of the present study should be acknowledged. First, the sample sizes in several of the incorporated studies were relatively small, which may reduce the strength of our conclusions. Second, all eligible studies were published in English and Chinese and indexed by the selected databases. It is possible that studies published in other languages or unpublished studies could be missed, which might bias the results. In addition, our result was on the basis of unadjusted estimates, while a more accurate analysis should be carried out if more detailed individual information was available, which would allow for an adjusted estimate by other causes such as age and sex. COX-2 polymorphisms, −765 G > C and +837 T > C, are linked to increased risk to oral cancers. However, the 1195 A > G polymorphism has no influence on oral cancer risk and will need to be explored further. More studies involving gene-gene and gene-environment interactions should also be taken into consideration in future analyses, which should lead to better, more comprehensive understanding of the correlation of the COX-2 gene polymorphisms with the risk of oral cancers.In summary, our meta-analysis revealed a strong association between the −765 G > C and +837 T > C polymorphisms and the susceptibility to oral cancer. Therefore,


CDKN2BAS gene and coronary heart disease (CHD). A total of 590 CHD cases and 482 non-CHD controls were involved in the present association study. A strong association of rs4977574 with CHD was observed in females  = 1.57, 95% confidential interval (CI) = 1.18–2.08). Moreover, rs4977574 was more likely to be a risk variant of CHD under the recessive model in females . A breakdown analysis by age had shown that there was an 87% increased risk of CHD for females younger than 65 years  = 2, p = 0.0002; allele: χ2 = 11.31, df = 1, p = 0.0008, OR = 1.87, 95% CI = 1.30–2.70). Similar observation was also found in males younger than 65 years . p values were adjusted by age, sex, smoking, high density lipoprotein cholesterol (HDL-C) and low density lipoprotein cholesterol (LDL-C). Meta-analysis of 23 studies among 36,452 cases and 39,781 controls showed a strong association between rs4977574 and the risk of CHD .The goal of our study was to explore the significant association between a non-protein coding single nucleotide polymorphism (SNP) rs4977574 of  Coronary heart disease (CHD) is the top cause of human death in the United States  and AsiaCDKN2BAS gene is a large antisense non–coding RNA, which is differentially expressed in a variety of tissues such as vascular endothelial cells and smooth coronary muscle cells CDKN2BAS diabetes . HoweverCDKN2BAS gene and CHD in Han Chinese. In addition, a meta-analysis of 11 studies among 36,452 cases and 39,781 controls is also performed to evaluate the contribution of rs4977574 of CDKN2BAS gene to the risk of CHD.The goal of our study is to test the association between rs4977574 of p = 0.0001) or male subjects (p = 0.003) were presented in CHD group. Smokers were more frequent in patients with CHD (p = 0.018). The high-density lipoprotein cholesterol (HDL-C) levels in the CHD cases were much lower than in the controls (p = 0.001). The low density lipoprotein cholesterol (LDL-C) concentration revealed a strong difference between cases and controls (p = 0.003). The allele frequency and genotype distributions of rs4977574 were listed in p > 0.05), our case-control study was limited in females. As shown in p = 0.002; allele: p = 0.002, odd ratio (OR) = 1.57, 95% confidential interval (CI) = 1.18–2.08). Moreover, rs4977574 was more likely to be a risk variant of CHD under the recessive model in females . In addition, we performed a breakdown comparison by age between cases and controls  = 2, p = 0.0002; allele: χ2 = 11.31, df = 1, p = 0.0008, OR = 1.87, 95% CI = 1.30–2.70). No significant difference was found in females aged 65 years or older (p > 0.05). Interestingly, the male controls in different age groups showed no deviation of HWE (p > 0.05). And the similar significant association of rs4977574 with CHD was found in the males younger than 65 years .The characteristics of study participants were provided in controls . A signip = 0.001, I2 = 54.2%), random-effect method was applied for the meta-analysis. As shown in CDKN2BAS gene and the risk of CHD . No visual publication bias was shown in the Funnel plot and the Egger regression plot (p (z) < 0.0001) and the Asian studies  < 0.0001). Subgroup study indicated that the ethnicity was likely to be the main source of heterogeneity , in contrast of moderate heterogeneity in the Asian studies .Our meta-analysis was involved with 23 case-control studies (including four groups in our study) among 36,452 cases and 39,781 controls ,36,37,38ion plot . The metogeneity . SpecifiCDKN2BAS rs4977574 polymorphism with the risk of CHD in Han Chinese. Our results showed that rs4977574 was significantly associated with CHD in females . Interestingly, this SNP was more likely to increase the risk of CHD under the recessive model in females . In addition, we also found that rs4977574 might have an 87% and 45% increased risk of CHD in females and males younger than 65 years, respectively. Our meta–analysis among 36,452 cases and 39,781 controls concluded that rs4977574 of CDKN2BAS gene contributed to the risk of CHD in both Asian and Caucasian populations, although there was substantial heterogeneity among the involved studies (I2 = 54.2%).The goal of our case-control study was to explore the significant association of CDKN2BAS . This region has been considered as the most widely and consistently replicated risk locus for CHD and MI [CDKN2BAS is unknown, but the CDKN2BAS transcript level shows bold correlation with the severity of atherosclerosis [CDKN2BAS gene expression mediates susceptibility to several important human diseases such as CHD and cancer [CDKN2BAS gene in blood [CDKN2BAS is a new susceptibility gene for the risk of CHD [SNP rs4977574 on chromosome 9p21.3 is located in D and MI ,39. The clerosis . The modd cancer . The 9p2in blood . Evidenck of CHD ,36,40.n = 254 vs. 482 in male cases). Multiple-center collaboration for a better coverage of the Han Chinese population in Ningbo city is warranted for the association test of this variant in the future.Sex differences have been proved in the prevalence and clinical outcomes of subclinical peripheral artery disease (PAD) , and theet al. [LRP8) were risk factor of CHD. The risk haplotype TACGC in LRP8 existed only in patients with familial and early onset CHD [et al. [ALMS1) gene polymorphisms were shown to be significantly associated with early-onset MI in both Japanese and Korean populations [CDKN2BAS exerts its effects on the risk of CHD.Early-onset CHD is more suitable for genetic analysis , becauseet al.  reportednset CHD . Monika  [et al.  confirmeulations . Some evulations ,58,59,60p < 0.0001, OR = 1.27, 95% CI = 1.22–1.31). This agrees with the previous meta-analysis [The sample size is comparatively small for the breakdown analysis by age and gender. Although the sample size may not be the optimal, our power calculation for male and female (<65 year) suggests that our study has a 78.2% and 92.1% power to detect a relative risk at a significant level of 0.05 which should be sufficient to describe a tendency to guide clinical practice. Since all the subgroups meet with HWE, we integrate the four subgroups into the current meta-analysis. For the meta-analysis among a total of 36,452 cases and 39,781 controls, our results show that rs4977574-G can increase 27% of CHD risk  and 482 non-CHD controls  were recruited between May 2008 and April 2012 from the Lihuili Hospital in Ningbo city of Zhejiang province, China. Each of the CHD cases had stenosis greater than 50% in one or more major coronary arteries , and judBlood samples were obtained after a 12 h overnight fast from subjects using 3.2% citrate sodium-treated tubes. Plasma levels of triglycerides (TG), total cholesterol (TC), and HDL-C, LDL-C were enzymatically measured using standard methods on an Olympus AU2700 automatic analyzer .Genomic DNA was isolated from whole blood by the conventional phenol/chloroform extraction method and was stored in 200 μL of TE buffer. DNA concentration was quantified using the Biophotometer plus  with the manufacturerʼs protocol. The primers for the Polymerase Chain Reaction (PCR) were as followed: forward primer, 5'-ACGTTGGATGGGACATCTTTTGTTAGAGTG-3'; reverse primer, 5'-ACGTTGGATGGTTTGCTTTCAGGGTACATC-3'; extension primer, 5'-CCCGACATCAAATGCATTCTATAGC-3'. DNA amplification was performed on the ABI Geneamp PCR System 9700 Dual 384-Well Sample Block Module . PCR cycling program included a 15 s initial denaturation stage at 94 °C, followed by 45 cycles of denaturation for 20 s at 94 °C, annealing for 30 s at 56 °C, and extension at 72 °C for 1 min, and a final extension for 3 min at 72 °C. Allele specific primer extension was performed on the Sequenom MassARRAY iPLEX platform according to the manufacturer’s instructions . To veriCDKN2BAS”, “CDKN2B–AS1” or “ANRIL”, “polymorphism” and “genetic association”. Interested information was selected after reading the full text articles. Other articles from the reference list on the retrieved and previous meta–analysis of this subject were evaluated based on the potential relevance. In addition, the authors of the retrieved papers were contacted directly for any additional and unpublished data. The inclusion criteria for the study were as follows: (1) case-control or cohort study; (2) the studies for which odds ratios (ORs) and 95% CIs were given, or could be calculated on the basis of genotype and allele frequencies. A random-effect model was applied when heterogeneity was detected (I2 > 50%) [We examined all studies by a search of the papers published on the electronic databases , Web of Science  and the Cochrane Library ) from 2009 to 2012. Various combinations of keywords were used to the following search terms, such as “coronary heart disease” or “coronary artery disease” or “myocardial infarction” combined with “2 > 50%) . Data exT-test was applied for the association of CHD with continuous variables including age, TG, TC, HDL-C and LDL-C. Pearson chi-square or Fisher exact test was used for categorical variables including sex, smoking, hypertension, diabetes and family history. Multivariable logistic regression analysis was used to detect association of the parametric and nonparametric phenotypes with genotypes, respectively. The departures of HWE of the genotypes were analyzed by the Arlequin program  [p value <0.05 was considered to be significant.zerland) . Comparizerland) . ORs andzerland) . The powzerland) . Meta-anzerland) . The pubzerland) . AccordiOur case-control study has identified a significant association of rs4977574 with the risk of CHD under a recessive inheritance model in females. Meta-analysis of 23 studies among 36,452 cases and 39,781 controls has established rs4977574 as a risk factor of CHD in multiple populations including Asians and Caucasians.


ACE) gene insertion/deletion (I/D) polymorphism with ACE activity and obstructive sleep apnea syndrome (OSAS) to see whether ACE activity is causally associated with OSAS. Literature search and data abstraction were done in duplicate. Sixteen articles including 2060 OSAS patients and 1878 controls were summarized. Overall, no significance was observed for the association of I/D polymorphism with OSAS, whereas carriers of II genotype  or I allele  had a lower level of serum ACE activity compared with DD genotype carriers, respectively. In subgroup analyses, carriers of II genotype were 3.806 times more likely to develop OSAS  in OSAS patients with hypertension, without heterogeneity. Mendelian randomization analysis indicated there was 37.4% (95% CI: 1.115 to 3.142) and 32.4% (1.106 to 2.845) increased risk of OSAS by a reduction of 1 U/L in ACE activity for the II genotype and I allele carriers versus DD genotype carriers, respectively. There was no observable publication bias. Collectively, genetically-reduced serum ACE activity might be a causal risk factor for OSAS.We meta-analytically summarized the associations of angiotensin converting enzyme ( ACE) is recognized as a promising candidate in susceptibility to OSAS.As the most common category of sleep apnea, obstructive sleep apnea syndrome (OSAS) is estimated to affect 4% to 5% of the general population135et al. reported that OSAS patients had lower ACE activity than healthy controlset al., who reported that ACE activity was significantly higher in OSAS patients than in healthy controlsACE gene, with the D/D, I/D and I/I genotypes respectively paralleling the high, middle and low levels of ACE activity10ACE gene I/D polymorphism as an instrument variable under the assumptions of Mendelian randomization by a meta-analysis in accordance with the guidelines of Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) (Please see the http://www.prisma-statement.org/statement).ACE is involved in catalyzing the conversion of angiotensin I into a biologically active peptide angiotensin II, a potent vasopressor and aldosterone-stimulator that regulates blood pressure and fluid-electrolyte balancehttp://www.wanfangdata.com.cn/, Chinese) and CNKI  were reviewed to search potential articles that tested the associations of ACE gene I/D polymorphism or plasma/serum ACE activity with OSAS susceptibility. The key words were “angiotensin converting enzyme” or “angiotensin I converting enzyme” or “ACE”, or “ACE1” in the ABSTRACT, along with “sleep apnea” or “sleep apnoea” or “sleep disorder” or “OSAS” or “OSAHS” in the TITLE. The last search was updated on April 4, 2015. Only articles published in English or Chinese language were considered in this study. In order to avoid possible missing hits, the reference lists of major original articles and reviews were checked manually.Four public electronic datasets, PubMed, Web of Science, Wanfang  and there was no disagreement.Articles were qualified if they tested the associations of The following data were abstracted from each qualified article by two authors (Lan He and Bin Wang) independently based on a predetermined protocol: the first author’s last name, year of publication, race, study design, source of healthy controls, diagnostic criteria for OSAS and apnea hypopnea index (AHI) cutoff, as well as sample size, age, percentage of male gender, body mass index (BMI), smoking, drinking, AHI cutoff, Epworth sleepiness score, hypertension and diabetes mellitus between OSAS patients and healthy controls.II vs. DD. Secondly, we suppose that the mean difference of ACE activity between II genotype and DD genotype is expressed as ΔP. Thirdly, under the assumptions of Mendelian randomization, ORGG vs. gg1/ΔP is regarded as an unbiased and unconfounded estimate of the OR of OSAS for a unit change in ACE activity12Mendelian randomization is emerging as a novel epidemiologic study design that utilizes measured variation in genes of known biology function to examine the causal impact of a modifiable exposure on disease susceptibility in genetic association studiesI2, quantifies the magnitude of heterogeneity as a percentile, with a higher value paralleling a higher probability of heterogeneity. In fact, I2 is a derivative of the Q statistic and it measures the proportion of variability that is due to heterogeneity rather than sampling error. Generally, the cutoff of I2 is set at 50% to define statistically significant heterogeneityACE gene I/D polymorphism and ACE activity with OSAS susceptibility were respectively quantified as odds ratio (OR) and weighted mean difference (WMD), with 95% confidence interval (95% CI) by the DerSimonian and Laird random-effects modelInconsistency index, 2 and BMI < 30 kg/m2), hypertension status (with and without hypertension) and sample size (≥300 subjects and <300 subjects) can explain heterogeneity. Moreover, meta-regression analyses were conducted on continuous covariates including age, gender, BMI, smoking, drinking to seek possible explanations for heterogeneity.Subgroup analyses were conducted to see whether differences in race (mainly Asian and White), language of publication (English and Chinese), study design (retrospective and prospective), source of controls , obesity .ACE gene I/D polymorphism and ACE activity with OSAS susceptibility. Only 16 articles (9 articles in English and 7 articles in Chinese) including 2060 OSAS patients and 1878 controls were left for final analysis after applying eligibility criteria91617181920212223242526272829A comprehensive review of public electronic datasets produced a total of 179 eligible articles (62 articles in English and 117 articles in Chinese) that tested the associations of ACE gene I/D polymorphism with OSAS susceptibility under allelic, homozygous genotypic and dominant models, and heterogeneity was significant , significance was detected in OSAS patients with BMI of less than 30 kg/m2 under only allelic model , with evident heterogeneity (I2 = 79.8%). In addition, four of 16 qualified articles that reported genotype data by hypertension status were analyzed separately. In OSAS patients complicated with hypertension, the significant associations of ACE gene I/D polymorphism with OSAS susceptibility were identified under three genetic models. For example, carriers of II genotype were 3.806 times more likely to develop OSAS , and there was no indication of significant heterogeneity (I2 = 48.2%) (For the association of = 48.2%) .ACE gene I/D polymorphism and OSAS susceptibility. Echoing from Begg’s funnel plots and Egger’s regression tests, there was no suggestive publication bias .Further, meta-regression analyses modeling age, gender, BMI, smoking and drinking showed no statistical significance for the association between ion bias . As suggOn the basis of above estimates, implementation of Mendelian randomization identified 37.4%  and 32.4%  increased risk of developing OSAS by a reduction of 1 U/L in serum ACE activity for the II genotype and I allele carriers versus the DD genotype carriers, respectively.ACE gene I/D polymorphism can predict the risk for OSAS patients complicated with hypertension, and more importantly, genetically-reduced serum ACE activity might be a causal risk factor for OSAS by Mendelian randomization technique. To the best of our knowledge, this meta-analysis represents the first to test the causal contribution of ACE activity in the pathogenesis of OSAS.Polling previous studies on the ACE-OSAS relation, our findings demonstrate that ACE gene ranks high on the list. In spite of many candidate association studies conducted in different ethnic groups, data in medical literature are inconsistent and even contradictory, and biological implication from statistical inference remains elusive. Lately, two meta-analyses found no relationship between ACE gene I/D polymorphism and OSAS susceptibility318923Among the promising candidate genes identified so far in susceptibility to OSAS, ACE gene I/D polymorphism and OSAS susceptibility. To produce more information, we subsequently conducted a wide range of subgroup analyses to seek possible sources of heterogeneity resulting from clinical and methodological aspects, and interestingly a positive signal emerged after restricting data to OSAS patients complicated with hypertension and normotensive healthy controls. It is well recognized that OSAS is a risk factor for the development of hypertension and approximately half of OSAS patients have hypertension36ACE gene underlying OSAS susceptibility in connection with hypertension is beyond the capability of this meta-analysis, it is rational to speculate that the predictive role of ACE genetic alterations is manifested in the coexistence of hypertension.In agreement with the previous two meta-analyses31ACE gene I/D polymorphism as an instrument variable. Factually, the selection of I/D polymorphism as a surrogate to infer causality between ACE activity and OSAS is biologically robust as this polymorphism was reported to interpret half the variance of serum ACE activity23The most noteworthy finding of this study is that our Mendelian randomization analysis demonstrated that genetically-reduced serum ACE activity might be a causal risk factor for OSAS with the aid of ACE gene was selected as a surrogate marker, and it is encouraged to bring other functional polymorphisms within or flanking ACE gene together to examine their joint impact on OSAS susceptibility and ACE activity. Moreover, as with a majority of meta-analyses, heterogeneity is a disturbing issue for both overall and subgroup analyses, and seeking other sources of between-study heterogeneity still remains a challengeable task. Further, all retrieved articles were published in English or Chinese language, leading to a possibility of selection bias. However, as viewed from Begg’s funnel plots and Egger’s tests, there was no indicative of publication bias, substantiating the robustness of our findings.There are several research limitations for this meta-analysis. The conclusion of this study was based on 16 independent articles and 3938 subjects, which might not be sufficient enough to quantify the effect estimate reliably. Only one polymorphism in ACE gene I/D polymorphism can predict the risk for OSAS complicated with hypertension, and more importantly, genetically-reduced serum ACE activity might be a causal risk factor for OSAS. Nevertheless, the results of this meta-analysis need further replication in other ethnic groups, and in-vivo and in-vitro functional studies are also required to unravel potential molecule mechanisms underlying the associations of ACE genetic defects and its serum activity with OSAS susceptibility.In conclusion, through a well-designed Mendelian randomization meta-analysis, our findings demonstrate that How to cite this article: He, L. et al. Genetically-reduced serum ACE activity might be a causal risk factor for obstructive sleep apnea syndrome: A meta-analysis. Sci. Rep.5, 15267; doi: 10.1038/srep15267 (2015).


There is growing evidence that indicates the presence of a prothrombotic state in atrial fibrillation (AF). However, the role of hemostatic markers in AF remains inconclusive.We conducted a meta-analysis of observational studies to evaluate the association between hemostatic markers and AF. A meta-regression was performed to explore potential sources of heterogeneity.A total of 59 studies met our inclusion criteria for the meta-analysis. For platelet activation, increased circulating platelet factor-4, β-thromboglobulin (BTG) and P-selectin were significantly higher in AF cases compared with controls ]: 1.72[0.96–2.49], 1.61[1.03–2.19] and 0.50[0.23–0.77], respectively). For coagulation activation, increased levels of plasma D-dimer, fibrinogen, thrombin-antithrombin, prothrombin fragment 1+2, and antithrombin-III were significantly associated with AF . For fibrinolytic function, tissue-type plasminogen activator and plasminogen activator inhibitor-1 were significantly increased in AF cases compared with controls  but the associations became nonsignificant after performing subgroup analysis by anticoagulants treatment status. For endothelial function, increased von Willebrand factor was significantly associated with AF ; however, no association was observed for soluble thrombomodulin .Increased circulating hemostatic factors  are significantly associated with AF. Future research is necessary to elucidate the precise mechanism of the prothrombotic state and how hemostatic markers promote thromboembolism in AF. The prevalence of atrial fibrillation (AF) is estimated to be 0.4% to 1.0% in the general population and is increasing . The patAn increasing number of studies have been performed to delineate hemostatic function in AF. The abnormalities observed in AF primarily include abnormal endothelial function, platelet and coagulation activation, and abnormal fibrinolysis. For coagulation and platelet activation, some studies have demonstrated that coagulation and platelet activation markers were elevated in AF –7. HowevIdentifying the specific hemostatic markers that are associated with AF is important for understanding AF etiology, development and prognosis. Therefore, we conducted a meta-analysis to evaluate the associations between hemostatic markers  and AF.2-antiplasmin”, “fibrinopeptide A”, “tissue-type plasminogen activator”, “urokinase-type plasminogen activator”, “plasminogen activator inhibitor”, “plasmin-antiplasmin complex”, “von Willebrand factor” or “soluble thrombomodulin”, and “atrial fibrillation”. We also manually searched journals and the reference lists of all retrieved articles. Additionally, relevant review articles were also cross-referenced. The literature retrieval was performed in duplication by two independent reviewers (N.W. and S. T.).This meta-analysis was performed according to the guidelines presented in the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) . We usedHuman studies, regardless of sample size, were included if they met the following criteria: (1) the study design was a case-control study (retrospective or nested case-control) or cohort study (retrospective or prospective cohort study) and (2) the study investigated the association between hemostatic markers  and AF.2-antiplasmin, plasmin-antiplasmin complex and fibrinopeptide A, were investigated in less than two studies and were also excluded from this meta-analysis.When multiple publications were based on the same or overlapping data, we used the most recent or largest population as recommended by Little et al . The numTwo reviewers (N.W. and S. T.) independently assessed the study quality using the primary criteria for nonrandomized studies described in the Newcastle-Ottawa scale . A “scorUsing standard data extraction forms, two reviewers (N.W. and Y.X.) extracted data from the relevant studies. We extracted publication information , participant characteristics , the type of AF and hemostatic marker levels (the mean and SD of each group). Any disagreements were resolved by consensus with a third reviewer (L.Z.).http://digitizer.sourceforge.net) to read the exact means and SDs from the figures. This program was able to read the exact values by digitizing the data points in an image file after manually setting the axis coordinates.If the study provided the medians and ranges instead of the means and SDs, we imputed the means and SDs using the method developed by Hozo et al . For stuI2 statistic, which describes the proportion of total variation across studies that is due to heterogeneity rather than chance [I2≥50% suggests that there is significant heterogeneity between studies. The SMD was pooled using a random effects model to manage heterogeneity when the heterogeneity was significant (I2 ≥50%); otherwise, a fixed effects model was applied. We evaluated potential publication bias using funnel plots and Begg’s tests [The standardized mean difference (SMD) was used as a summary statistic in the meta-analysis. SMD is the difference in mean outcomes between groups divided by the standard deviation of the outcome among participants. The SMD method is commonly used when a variety of methods and units are used to measure hemostatic marker levels among different studies. In this circumstance, it was necessary to standardize the results of the studies to a uniform scale before they were able to be combined. The statistical heterogeneity across studies was assessed using the n chance . An I2≥’s tests .P<0.05) in univariate model. A subgroup analysis was performed by categorical variables which were significant in the univariate meta-regression model to explore how these factors influenced the level of hemostatic markers. Considering anticoagulant treatment had great impact on levels of hemostatic markers, we also carried out a subgroup analysis by anticoagulants treatment status as follows: Subgroup 1, there was significant difference in the proportion of subjects received anticoagulants between AF patients and controls; Subgroup 2, no patient received anticoagulants in both groups; Subgroup 3, all participants were anticoagulated in both groups; Subgroup 4, there was no significant difference in the proportion of anticoagulation treatment in both groups (excluding the Subgroup 2 and 3); Subgroup 5, anticoagulation information was not available in both groups. The significance level for all analyses was set at P<0.05. Statistical analyses were performed using the STATA software package .A weighted random-effect meta-regression analysis was performed to examine potential sources of heterogeneity and potential confounding factors. The weight for each trial was equal to the inverse of the sum of the within trial variance and the residual between trial variance. The restricted maximum likelihood (REML) method was used to estimate the residual between trial variance. The true effect of each included study was treated as random-effect, and the true effects of included studies are a random sample of the relevant distribution of true effects . Based oA total of 661 records were retrieved after the primary literature search. After screening the titles and abstracts, 531 studies were excluded; 106 potentially relevant full-text articles were reviewed, and 59 studies were included in the meta-analysis . Of the P = 0.066) (P = 0.025), 1.72 , 1.61 , and 0.50 , respectively  . Mean plectively –2E.2 above 90% (2 = 49.3% (P = 0.139) and two permanent AF studies with an I2 = 0.0% (P = 0.895)  . The res= 0.895) .P<0.001), 0.72 , 0.42 , 1.00 , and 1.38 , respectively , prothrombin fragment 1+2 (F1+2), and antithrombin- III (AT- III), were significantly higher in patients with AF than controls, with a pooled SMD of 1.82  and 0.87 , respectively  and plasminogen activator inhibitor-1 (PAI-1). A meta-analysis of 9 studies for tPA and 11 studies for PAI-1 determined that both markers were significantly increased in AF cases compared with controls, with a pooled SMD of 0.86  (P = 0.107) (P = 0.0505)  was significantly elevated in AF patients compared with controls, with a pooled SMD of 0.79  . No sign= 0.107) . A univa= 0.107) . However 0.0505) . A subgr 0.0505) . The res 0.0505) .This meta-analysis summarized the evidence to date regarding the association between a series of hemostatic markers and AF. Our results demonstrated that increased circulating PF-4, BTG, P-selectin, D-dimer, fibrinogen, TAT, F1+2, AT- III, and vWf levels were significantly associated with AF.Platelet and coagulation activation, fibrinolytic dysfunction, and endothelial damage are four major components of the hemostatic mechanism. Several studies have demonstrated that certain hemostatic markers are associated with AF, but these associations remain controversial , 73. OurThe exact mechanism of the association between the abnormal levels of hemostatic markers and AF is unclear. Whether abnormal level of hemostatic markers is a consequence of AF or the presence of alteration in hemostatic markers promotes AF development is still under debate. Accumulative proofs have implied that both mechanisms may interrelate, which means that hemostatic markers are not only a consequence but also a cause. Lip proposed that AF can confer a hypercoagulable state as early as 1995 . ResearcAdditionally, hemostatic markers can be altered by antithrombotic medications. A study has demonstrated that after receiving anticoagulation, the levels of D-dimer and F1+2 in AF patients were lowered . TherefoAF is sometimes complicated by other cardiovascular diseases, and hemostatic markers are known to be influenced by these comorbidities. Whether hemostatic markers are increased due to AF itself or are coexisting cardiovascular risk factors remains a matter of debate . Feng etThere is also a strong association between AF and the risk of thromboembolic stroke, although the mechanism for this increased risk has not been fully determined. The pathogenesis is multifactorial and stasis in a poorly contractile left atrium is partially responsible . In addiTo the best of our knowledge, this is the first systematic and comprehensive meta-analysis examining the associations between a series of hemostatic markers and AF. Due to the currently high incidence of AF and subsequent thromboembolic events, identifying of biological markers that can predict AF risk and hypercoagulable propensity has important clinical implications. However, several limitations of this study should be considered when interpreting our results. High heterogeneity exists among studies. Therefore, we performed a meta-regression analysis to identify potential sources of heterogeneity; gender, type of AF, and comorbidities may explain a portion of this heterogeneity. However, not all of the relevant covariates, such as heart failure and inflammatory factors, are available for each study included in our meta-regression. Therefore, the heterogeneity and unknown confounders may still influence true associations. Secondly, observational studies especially case-control studies could only determine whether there was an excess risk of adverse outcome in association with biomarkers. Verifying causality requires more high-quality, large sample, and more stringent evidence. Our study as a systematic review and meta-analysis summarized the evidences in this field and could provide useful clues and generate initial data for future studies. Finally, statistical tests suggested that there might be publication bias for studies examining BTG, D-dimer, fibrinogen, AT- III and vWf .In conclusion, our meta-analysis demonstrated that elevated circulating hemostatic markers  were associated with AF. Our findings provide useful clues and evidences for future study to elucidate the precise mechanism of the prothrombotic state and the role of hemostatic markers in promoting thromboembolism in AF patients.S1 Checklist(DOC)Click here for additional data file.S1 FigBlack diamond indicates the SMD, with the size of the square inversely proportional to its variance, and horizontal lines represent 95% CI. The pooled results are indicated by the black hollow diamond. AF, atrial fibrillation; PAF, paroxysmal AF; PtAF, permanent AF; SMD, standardized mean difference; CI, confidence interval.(JPG)Click here for additional data file.S2 FigBlack diamond indicates the SMD, with the size of the square inversely proportional to its variance, and horizontal lines represent 95% CI. The pooled results are indicated by the black hollow diamond. AF, atrial fibrillation; MPV, mean platelet volume; SMD, standardized mean difference; CI, confidence interval; Subgroup 1, there was significant difference in the proportion of subjects received anticoagulants between AF patients and controls; Subgroup 4, there was no significant difference in the proportion of anticoagulation treatment in both groups .(JPG)Click here for additional data file.S3 FigBlack diamond indicates the SMD, with the size of the square inversely proportional to its variance, and horizontal lines represent 95% CI. The pooled results are indicated by the black hollow diamond. AF, atrial fibrillation; PAF, paroxysmal AF; PtAF, permanent AF; SMD, standardized mean difference; CI, confidence interval.(JPG)Click here for additional data file.S4 FigBlack diamond indicates the SMD, with the size of the square inversely proportional to its variance, and horizontal lines represent 95% CI. The pooled results are indicated by the black hollow diamond. AF, atrial fibrillation; > = 2000, publication year after 2000; <2000, publication year before 2000; SMD, standardized mean difference; CI, confidence interval.(JPG)Click here for additional data file.S5 FigBlack diamond indicates the SMD, with the size of the square inversely proportional to its variance, and horizontal lines represent 95% CI. The pooled results are indicated by the black hollow diamond. AF, atrial fibrillation; tPA, tissue-type plasminogen activator; SMD, standardized mean difference; CI, confidence interval; Subgroup 1, there was significant difference in the proportion of subjects received anticoagulants between AF patients and controls; Subgroup 2, no patient received anticoagulants in both groups; Subgroup 5, anticoagulation information was not available in both groups.(JPG)Click here for additional data file.S6 FigBlack diamond indicates the SMD, with the size of the square inversely proportional to its variance, and horizontal lines represent 95% CI. The pooled results are indicated by the black hollow diamond. AF, atrial fibrillation; PAI-1, plasminogen activator inhibitor-1; SMD, standardized mean difference; CI, confidence interval; Subgroup 1, there was significant difference in the proportion of subjects received anticoagulants between AF patients and controls; Subgroup 2, no patient received anticoagulants in both groups; Subgroup 5, anticoagulation information was not available in both groups.(JPG)Click here for additional data file.S7 FigBlack diamond indicates the SMD, with the size of the square inversely proportional to its variance, and horizontal lines represent 95% CI. The pooled results are indicated by the black hollow diamond. vWf, vonWillebrand factor; AF, atrial fibrillation; PAF, paroxysmal AF; PeAF, persistent AF: PtAF, permanent AF; aAF, acute AF; CAF, chronic AF; SMD, standardized mean difference; CI, confidence interval.(JPG)Click here for additional data file.S8 FigBlack diamond indicates the SMD, with the size of the square inversely proportional to its variance, and horizontal lines represent 95% CI. The pooled results are indicated by the black hollow diamond. AF, atrial fibrillation; PAF, paroxysmal AF; PeAF, persistent AF: PtAF, permanent AF; aAF, acute AF; CAF, chronic AF; SMD, standardized mean difference; CI, confidence interval; Subgroup 1, there was significant difference in the proportion of subjects received anticoagulants between AF patients and controls; Subgroup 2, no patient received anticoagulants in both groups; Subgroup 4, there was no significant difference in the proportion of anticoagulation treatment in both groups; Subgroup 5, anticoagulation information was not available in both groups.(JPG)Click here for additional data file.S9 FigP values were calculated by Begg’s test. A. Platelet count and AF (P = 0.484); B. MPV and AF (P = 0.462); C. PF-4 and AF (P = 0.584); D. BTG and AF (P = 0.042); E. P-selectin and AF (P = 0.535); F. D-dimer and AF (P = 0.001); G. Fibrinogen and AF (P<0.001); H. TAT and AF (P = 1.000); I. F1+2 and AF (P = 0.466); J. AT-III and AF (P = 0.024); K. tPA and AF (P = 1.000); L. PAI-1 and AF (P = 0.661); M. vWf and AF (P<0.001); N. sTM and AF (P = 0.452).Each point represents a separate study for the indicated association. (DOC)Click here for additional data file.S1 Table(DOC)Click here for additional data file.S2 Table(DOC)Click here for additional data file.S3 Table(DOC)Click here for additional data file.S4 Table(DOC)Click here for additional data file.


BTNL2) rs2076530 gene polymorphism has been implicated in susceptibility to sarcoidosis. However, results from previous studies are not consistent. To assess the association of BTNL2 polymorphism and sarcoidosis susceptibility, a meta-analysis was performed.Butyrophilin-like 2  with 95% confidence intervals (CI) were calculated.BTNL2 rs2076530 polymorphism was associated with sarcoidosis susceptibility in allelic model , dominant model , and recessive model .Ten studies involving a total of 3303 cases and 2514 controls were included in this meta-analysis. Combined data indicated that BTNL2 rs2076530 polymorphism contributes to the risk of sarcoidosis.This meta-analysis indicates that  BTNL2) gene.Sarcoidosis is a multisystemic inflammatory disease with unknown aetiology, characterized by the formation of noncaseating granulomas and an exaggerated cellular immune response due to increased inflammatory activity of macrophages and CD4 helper T cells,2. Many BTNL2, a member of the immunoglobulin superfamily, reduces proliferation and cytokine production from activated T cells, suggesting a role of functional BTNL2 as a negative costimulatory molecule involved in T-cell activation[BTNL2 seems to be involved in sarcoidosis. Several studies have been undertaken to evaluate this potential relationship between BTNL2 rs2076530 polymorphism and sarcoidosis susceptibility. However, results from different reports of different geographic areas are not consistent. Besides, many studies were small in size and, as such, may on their own lack sufficient statistical power to address this issue adequately. In an effort to clarify the association of BTNL2 rs2076530 polymorphism and sarcoidosis susceptibility, we performed a meta-analysis on all the studies identified by systematic review of literatures.ctivation. The G→Activation,5. So thThis meta-analysis followed the Preferred Reporting Items for Systematic Reviews and Meta-analysis (PRISMA) criteria. Two invBTNL2 rs2076530 polymorphism and sarcoidosis risks; (2) case-control designs on unrelated individuals; (3) odds ratios (OR) with 95% confidence intervals (CI) or sufficient data for their calculation were available; (4) the control subjects satisfied the Hardy-Weinberg equilibrium (HWE); (5) the diagnoses of sarcoidosis were consistent with the International Consensus Statement on Sarcoidosis[Reports that fulfilled the following criteria were included: (1) evaluating the rcoidosis or verifTwo authors evaluated the study quality independently according to the Newcastle-Ottawa scale (NOS), which aThe data were extracted by two investigators  independently and a consensus was reached on all items. Any disagreement was resolved as described above. The following data was collected: first author, year, ethnicity, sample size, mean age, gender ratio, sarcoidosis criteria, source of controls, and genotype or allele distribution in cases and controls.2 test was used to determine whether the observed frequencies of genotypes in control groups conformed to HWE. The strength of the associations between BTNL2 rs2076530 polymorphism and sarcoidosis risk was evaluated using OR with 95% CI. The pooled ORs were calculated for allelic model (A vs. G), dominant model (AA + AG vs. GG) and recessive model (AA vs. AG + GG) respectively. Heterogeneity was calculated using the Q-test and I2-statistics[P-value of the Q-test was>0.10, the pooled OR was assessed in a fixed-effects model; otherwise, the random-effects model was applied[I2 values were used to quantify heterogeneity[Z-test were used to determine the significance of the pooled OR, in which a P-value of <0.05 was considered statistically significant. Subgroup analyses were performed to evaluate the ethnic-specific effects. For the subgroup analysis by ethnicity, the study populations were stratified into three groups: Caucasians, Asians, and mixed population. We also performed sensitivity analysis by extracting a single study each time to check the stability of the results. Publication bias was assessed using Egger’s test, Begg’s test and funnel plot.The data were analyzed using Review Manager 5.2  and STATA 12.0 software . Pearson’s χtatistics. If the s applied. I2 valBTNL2 rs2076530 polymorphism and sarcoidosis susceptibility. After reading the full text, two articles were excluded for not enough data available. Finally, ten articles were included (the subjects of eight studies were Caucasion population[BTNL2 gene rs2076530 in sarcoidosis group (containing 1629 subjects) and control group (containing 1065 subjects). In the other five articles[The procedure for including eligible studies is shown in opulation,17,18,19opulation), contaiopulation,16,–21, opulation,16,19,21 articles,17,18,19 articles was perfBTNL2 rs2076530 was associated with an increased risk of sarcoidosis in allelic model . The pooled OR was 1.59 (95%CI: 1.47–1.72) under the fixed effect model, without between-study heterogeneity  (P<0.00001), Asian , and mixed population   . In subg0.00001) .vs. GG, OR = 2.10, 95%CI: 1.67–2.65, P<0.00001, vs. AG + GG, OR = 1.93, 95%CI: 1.49–2.50, P <0.00001, Combined data from the five studies with available genotype data , 19,21 sBTNL2 rs2076530 polymorphism and sarcoidosis susceptibility among the overall populations, the observed significant result was not materially altered after sequentially excluding each study.We performed sensitivity analysis for statistically significant result. For the association of the P = 0.956 and P = 0.592, respectively) in allelic model. The shape of the funnel plot was symmetrical Click here for additional data file.S2 File(DOC)Click here for additional data file.S3 File(DOCX)Click here for additional data file.S4 File(DOCX)Click here for additional data file.


Many observational studies have found that exposure to dental X-rays is associated with the risk of development of meningioma. However, these findings are inconsistent. We conducted a meta-analysis to assess the relationship between exposure to dental X-rays and the risk of development of meningioma.The PubMed and EMBASE databases were searched to identify eligible studies. Summary odds ratio (OR) estimates and 95% confidence intervals (95% CIs) were used to compute the risk of meningioma development according to heterogeneity. Subgroup and sensitivity analyses were performed to further explore the potential heterogeneity. Finally, publication bias was assessed.Seven case-control studies involving 6,174 patients and 19,459 controls were included in the meta-analysis. Neither exposure to dental X-rays nor performance of full-mouth panorex X-rays was associated with an increased risk of development of meningioma . However, exposure to bitewing X-rays was associated with a slightly increased risk of development of meningioma . Similar results were obtained in the subgroup and sensitivity analyses. Little evidence of publication bias was observed.Based on the currently limited data, there is no association between exposure to dental X-rays and the risk of development of meningioma. However, these results should be cautiously interpreted because of the heterogeneity among studies. Additional large, high-quality clinical trials are needed to evaluate the association between exposure to dental X-rays and the risk of development of meningioma. Intracranial meningioma is a solid tumor that arises from the meninges, which protect the central nervous system. Meningioma, one of the most clinically serious tumors of the central nervous system, accounts for approximately 20% of all intracranial tumors in male patients and 38% in female patients –4. The ePrevious studies have shown that biological risk factors of ionizing radiation (IR) are associated with a high incidence of atomic bomb-induced health issues, including various types of cancer , 7. ExpoWe searched the PubMed and EMBASE databases up to March 2014 to identify relevant studies that evaluated the association between exposure to dental X-ray and the risk of development of meningioma. The following search terms were employed: (1) “meningioma(s),” “brain neoplasm(s),” “brain tumor(s),” “brain Neoplasm(s),” “meningeal Neoplasm(s),” and (2) “dental x-ray(s),” “tooth radiography,” “teeth radiography,” and “dental radiography.” Furthermore, we reviewed the reference lists of all eligible articles.Studies were considered eligible for inclusion in the meta-analysis if they met the following criteria: they evaluated the association between exposure to dental, full-mouth panorex, and bitewing X-ray exposure and the risk of development of meningioma; the full text of the article was available; and they provided the adjusted and/or unadjusted odds ratio (OR), relative risk (RR) with corresponding 95% confidence interval (95%CI), or raw data with which to estimate the crude OR or RR. Studies were excluded if they met the following criteria: they were letters, comments, correspondence, conference reports, or laboratory studies or they did not contain enough data with which to calculate the OR. When multiple publications covered the same study population, only the study with the larger sample was included. Two authors (J.K.S. and X.H.Y.) independently assessed the inclusion of all retrieved studies and resolved any disagreements through discussion or consultation with a third author (G.L.H.).Two authors (J.K.S. and X.H.Y.) independently collected the following basic information using a standardized data extraction form: first author’s name, year of publication, study design, source of control, study location, number of participants (cases/controls), crude and/or adjusted point estimates and corresponding 95%CIs, and covariate features included in the multivariable model. Disputes were resolved by discussion and consensus with a third author (G.L.H.).We evaluated the methodological quality of each study using the Newcastle-Ottawa scale (NOS) . Three m2 statistic of >50% was considered statistically significant. A random-effects model was used if heterogeneity was observed, while a fixed-effects model was used if the P value was >0.05 and/or the I2 statistic was <50%. We further performed sensitivity analyses to evaluate robustness and stability by sequential omission of one study in each turn. Moreover, subgroup analyses were performed to explore the potential presence of heterogeneity and assess the influence of different inclusion criteria on the overall estimate. Publication bias was evaluated using the Begg and Egger tests [We used the OR with 95%CI as a common measure across all eligible studies. Because meningioma is a relatively rare disease, differences among the estimates of relative risk were ignored and the RR was directly converted to the OR. We used the Cochrane Q test to evaluate statistical heterogeneity. A P value of <0.05 and/or an Ier tests , 19. AllThe characteristics of all included articles are presented in All seven studies reported exposure to dental X-rays; four reported exposure to full-mouth series , 13, 16,We used the NOS to evaluate the quality of the included studies . The med2 = 86.5%). The results suggested that exposure to dental X-rays has no important effects on the risk of development of meningioma, and substantial heterogeneity was observed (2 = 48.2%). The subgroup analyses based on different exclusion criteria yielded similar results  with significant heterogeneity . These results suggested that exposure to dental X-rays has no important effects on the risk of development of meningioma. No significant heterogeneity was observed . These results suggest that exposure to dental bitewing X-rays is associated with a slightly increased risk of development of meningioma. Substantial heterogeneity was observed   .Both the Begg rank correlation test and the Egger funnel plot asymmetry test (regression method) in the meta-analysis indicated no significant publication bias  . BecauseThe pathogenetic mechanism of meningioma remains unknown. Many conditions have been identified as risk factors, including brain injury, smoking, chronic virus infection, and occupational exposure. The most threatening risk factor for the development of meningioma is exposure to IR. Notably, a moderate to high dose of IR could increase the risk of developing various cancers as confirmed by studies of atomic bomb survivors and children irradiated for benign medical conditions and primary tumors. However, most of the general population receives lower-dose exposure during procedures such as diagnostic radiography, computed tomography (CT), or other types of radiation. The precise nature of the relationship between exposure to IR  and the development of meningioma is not well characterized. With the progression of medical technology, medical diagnostic X-ray instruments have become widely used among the general population. Dental X-rays have been widely employed since 1919 . Four raThe meta-analysis design serves as a valuable tool with which to study the rare effects of an intervention or treatment, permitting data synthesis and providing more convincing estimates of effect. To the best of our knowledge, this is the first meta-analysis to explore the role of exposure to dental X-rays in patients with meningioma. The overall results of the present meta-analysis of seven case-control studies using a random-effects model provide evidence that exposure to dental X-rays is not likely to have any important effects on the risk of development of meningioma. The pooled estimates were robust across the sensitivity analyses, and no publication bias was observed. Exposure to dental full-mouth and panorex X-rays may not increase the risk of development of meningioma, while exposure to dental bitewing X-rays may slightly increase the risk. These findings can partly explained as follows. First, only two studies reported an association between exposure to dental bitewing X-rays and the risk of development of meningioma. Second, patients who undergo dental bitewing examination are exposed to a relatively higher dose of X-rays. However, because of the few studies that reported this association, the present meta-analysis does not have enough power for a decisive conclusion. Further studies should focus on this association. Only three studies reported the frequency of dental X-rays, and all showed negative associations. Because the classification of exposure differed among the studies, it is difficult to merge the results using a meta-analysis.Some limitations should be considered in the present meta-analysis. First, this meta-analysis was based on case-control studies. Although the case-control study is the most appropriate design for exposure causing rare event, this design has inherent limitations such as selective bias and recall or memory bias. Additionally, some confounding factors  are difficult to control in case-control studies. Second, substantial heterogeneity was a potential problem when interpreting the results of our analysis. This heterogeneity was not unexpected considering the differences in the characteristics of the study designs, populations sources, methods of information collection, and methods of ascertainment of exposure and outcomes among the included studies. After revealing the substantial heterogeneity, we performed subgroup analyses and obtained similar findings. Furthermore, sensitivity analyses were conducted and yielded similar results. When the study conducted by Lin et al.  was exclThe following points should be considered for further studies. First, it is necessary to establish a standardized protocol with respect to exposure dose, method of examination, and duration of exposure to dental X-rays because great variation exists in the literature. Second, more large-scale studies should be performed on the relationship between various types of dental X-ray exposure and the risk of development of meningioma. Finally, the average follow-up period was not reported in this meta-analysis; longer-term studies are needed.In conclusion, the currently available evidence indicates that exposure to dental X-rays is unlikely to have any important effects on the risk of development of meningioma. Although these findings are encouraging, the results of this meta-analysis should be interpret with caution because of the heterogeneity among the studies and the relatively limited number of studies, Further large-scale, well-designed trials on this topic are needed.S1 PRISMA Checklist(DOC)Click here for additional data file.


Two common polymorphisms in the IKZF1 gene (rs4132601 and rs11978267 variants) have been reported to be associated with childhood acute leukemia (AL) risk, however the results were inconsistent. Here, we conducted a meta-analysis to generate large-scale evidence on whether IKZF1 variants are risk factors for childhood AL.The PubMed, Embase, EBSCO, and Web of Science were searched up to June 2, 2014 for studies on the association of IKZF1 polymorphisms with childhood AL risk. Data were extracted and the odd ratios (ORs) and95% confidence intervals (95% CIs) were calculated by a fixed-effects orrandom-effects model. Subgroup analysis by ethnicity and leukemia subtype, sensitivity and cumulative meta-analyses were performed. Moreover, publication bias was assessed by Begg's and Egger's tests.In total, 33 case control studies were finally included in this meta-analysis. For rs4132601 polymorphism, significantly increased AL risk was observed in all genetic models . In the subgroup analysis by tumor type, statistical association was observed in B-cell precursor ALL (BCP-ALL). Additionally, when stratified by ethnicity, significantly increased AL risk was only observed in European subgroup, but not among African or mixed population subgroups. Finally, similar results were found forrs11978267 polymorphism.In summary, this meta-analysis provides evidence that rs4132601 and rs11978267 polymorphisms in the IKZF1 gene mightcontribute to the occurrence of BCP-ALL, especially in European populations. Moreover, further studies with large sample size are required to clarify possibleroles of IKZF1 variants in other ethnic groups . Acute leukemia, the most common type of childhood cancer and the leading cause of cancer-related deaths among children, affects 35–50 per 1,000,000 children per year Recently, accumulating studies suggest that inherited genetic factors affect the risk of developing ALL. Two genome-wide association (GWA) studies have identified SNPs in 7p12.2 (IKZF1), 9p21 (CDKN2A), 10q21.2 (ARID5B), and 14q11.2 (CEBPE)that contribute to susceptibility to ALL To date, one meta-analysis focused on the correlation between IKZF1 variants and ALL risk, which only investigated the association of one polymorphism (rs4132601) and ALL risk in the overall population. Moreover, some studies involving childhood acute leukemia were not included We systematically searched PubMed, Embase, EBSCO, and Web of Scienceusing the following search terms:, and . The search was last performed in June 2014. Moreover, the reference lists of retrieved articles were checked for additional potential studies.A study was eligible in the meta-analysis if it: (1) investigated the association of IKZF1 polymorphisms with childhood acute leukemia susceptibility (2) provided sufficient data on allele or genotype distribution in patients and controls. The exclusion criteria were: (1) no control population (2) the subjects of the study were adults (3) comments, review articles, meta-analysis, or articles only with an abstract.From each study, the following data was extracted independently by two authors: first author, publication year, country and ethnicity of the subjects, gender component, mean age of the study subjects, genotyping method, number of patients and controls, types of acute leukemia, allele and genotype frequency of patients and controls. In addition, if the genotype distribution was unavailable in the article, the corresponding author was contacted for the detailed data. Disagreements were resolved by discussion between the two investigators.The quality of each study was independently assessed by 2 authors using the quality scoring scale modified from previous meta-analysis of genetic studies The strength of the association between IKZF1 polymorphisms (rs4132601 or rs11978267) and childhood acute leukemia was measured by odds ratios (ORs) and corresponding 95% confidence intervals (CIs). The significance of the pooled OR was determined by the Z-test, and the P values were adjusted using Bonferroni correction by the number of compared SNPs (p = 0.05/2 = 0.025).Stratified analysis was performed according to types of AL (B-cell precursor ALL (BCP-ALL), T-cell ALL and AML) and ethnicity . Additionally, the Hardy-Weinberg equilibrium (HWE) of the control group was assessed, and a P value of less than 0.05 was considered significant disequilibrium.2-based Q test and I2 test, and heterogeneity was considered significant when a P value was less than 0.10 2 in meta-regression is the residual between-study variance that describes the variation in the results that is not explained by the covariates Heterogeneity across studies was assessed by χSensitivity analysis was performed by sequentially omitting one study each time to assess the effect of a single study on the pooled ORs. In addition, cumulative meta-analyses were also carried out for both variants in association with AL to evaluate the trend of the genetic risk effect (OR) of the allele contrast as evidence accumulating over time. Finally, publication bias was assessed using graphical evaluation of Begg's funnel plots and the Egger's regression test, a p value of less than 0.05 was considered as significant The combined search yielded 165 references from PubMed, Embase, EBSCO, and Web of Science databases. After review of titles and abstracts, 141non-relevant articles were excluded, including review articles, meta-analysis, articles only with an abstract, and duplicate studies. Full texts of the remaining 24 articles were reviewed and analyzed in detail, of which, 2 articles reported in adults, 2 investigated other variants in IKZF1 gene, and 5 did not have sufficient data. Finally, a total of 15 relevant articles involving the associations between polymorphisms in IKZF1 and risk of childhood AL were eligible for this meta-analysis The association between rs4132601 polymorphism and susceptibility to AL was analyzed in21 studies involving 5823 AL patients and 11393 healthy controls. Overall, the results of combined analyses showed a significantly increased risk of AL in all genetic models.   In the sA total of 20 studies with 4960 patients and 28034 controls were eligible for the pooled analysis of rs11978267 polymorphism. Meta-analysis findings of associations between rs11978267 polymorphism in IKZF1 gene and susceptibility of acute leukemia were summarized in 2 in the analysis of rs11978267 and rs4132601 polymorphism, respectively. In addition, the heterogeneity was removed in T-cell ALL or Asians subgroup of rs4132601 variant and in BCP-ALL, T-cell ALL or mixed population subgroup of rs11978267 variant.  For rs11978267 variant, however, significant publication bias was detected in most comparisons (GG vs TT: p = 0.01). IKZF1, located on chromosome 7p12.2,is an essential regulator of lymphopoiesis and plays an important role in the development of lymphoid lineages, especially in the commitment of CD4 and CD8 T-cell lineages In this meta-analysis, we observed a significantly increased AL risk in the analysis of rs4132601 polymorphism in IKZF1 gene in all genetic contrasts. When the data were stratified by disease type, a significant association was found in BCP-ALL subgroup, but not among T-cell ALL or AML subgroups. Similar results were found with thers11978267 variant. It is widely accepted that childhood ALisnot a single homogeneous disease and canbeclassified into subtypes: acute lymphoblastic (ALL) and myeloid leukaemia (AML), eachwith their own characteristics and potentially different aetiologies In the subgroup analysis by ethnicity, statistical correlation was observed in Europeans for both variants. However, no significantly increased ALL risk was found in Asians forrs4132601 polymorphism and African or mixed populations for rs11978267 polymorphism, suggesting that the relative contribution of susceptibility genes may vary across different ethnicities. International variation in the incidence of leukaemia, especially ALL, is well recognized, which was 44% higher among Whites compared to Blacks  2 in the analysis of rs11978267 polymorphism. Moreover, the heterogeneity was removed in BCP-ALL, T-cell ALL and AML subgroups under heterozygote comparison.  Furthermore, Galbraith plot analyses was also carried out to visualize the impact of individual studies on the overall heterogeneity, which indicated that 5 studies were the possible origin of heterogeneity, when excluded, the heterogeneity was removed and the association was still significant.  In addition, sensitivity analysis showed that no single study qualitatively changed the pooled odds ratios. Also, results of cumulative meta-analysis showed the pooled ORs tended to be stable and the associations tended toward significant as more data accumulating over time, indicating that the results of this meta-analysis are stable.Significant between-study heterogeneity existed when all 27 studies were pooled. We found the heterogeneity was remarkably decreased or even removed among Asian or mixed population subgroups, T-cell ALL subgroup and BCP-ALL subgroup. We then performed meta-regression to explore the potential source contributing to the heterogeneity, which suggested that types of disease might be a potential source of heterogeneity. The results indicated that disease type could explain 86.7% of τSeveral limitations of our study should be acknowledged. First, only1 and 5 studies were included in Africans and Asians, respectively. Thus, the association between IKZF1 polymorphisms and childhood acute leukemia in different populations need to be validated by further studies with large sample size. Second, a language bias may have occurred because of only studies published in English were included. Also, significant heterogeneity between studies was detected in this meta-analysis. Finally, the etiology of childhood ALL is believed to be multi-factorial, including genetic variables, infections, and environmental risk factors such as ionizing radiation. However, we could not perform gene-environment or gene-gene interactions due to the insufficient data.In summary, despite these limitations, our results were still significant. The results of this meta-analysis suggested that rs4132601 and rs11978267 polymorphisms in IKZF1 gene might contribute to the occurrence of BCP-ALL, especially in European populations. However, the association in other ethnic groups  needs to be validated in further studies with large sample size. Moreover, studies involving gene-gene and gene-environment interactions are required to clarify possible roles of multiple risk factors in childhood AL.Figure S1Sensitivity analysis on the associations between IKZF1 rs11978267 variant and childhood AL risk in allelic contrast model (G vs A). Results were computed by omitting each study (left column) in turn, Bars: 95% confidence interval.(TIF)Click here for additional data file.Figure S2Cumulative meta-analysis: pooled OR with the corresponding 95% CI at the end of each year information step is shown for IKZF1 rs11978267 polymorphism in allelic contrast model (G vs A). CI: confidence interval; OR: Odds ratio.(TIF)Click here for additional data file.Figure S3Galbraith plots of IKZF1 rs4132601 (A) or rs11978267 (B) polymorphism and childhood AL risk, which indicated the outliers as possible sources of heterogeneity. The regression runs through the origin interval . The 95% confidence interval is between the two outer parallel lines at two units above and below the regression line.(TIF)Click here for additional data file.Table S1Scale for quality assessment.(DOC)Click here for additional data file.Table S2P value for the Egger's test in the analysis of publication bias.(DOC)Click here for additional data file.Checklist S1PRISMA checklist.(DOC)Click here for additional data file.


The prevalence of telomerase reverse transcriptase (TERT) promoter mutations (pTERTm) in non-small-cell lung cancer (NSCLC) have been investigated, but the results were inconsistent. In addition, several studies have analysed the role of pTERTm in the etiology of various types of cancers, however, the results also remain inconsistent.The genomic DNA sequence of 103 NSCLC samples were analysed to investigate the frequency of pTERTm in these patients and to establish whether these mutations are associated with their clinical data. Furthermore, a meta-analysis based on previously published articles and our cohort study was performed to investigate the association of pTERTm with patient gender, age at diagnosis, metastasis status, tumour stage and cancer prognosis .In the cohort study, 4 patients had C228T and 2 had C250T, with a total mutation frequency up to 5.8%. Significant difference of clinical data between pTERTm carriers and noncarriers was only found in age at diagnosis. In the meta-analysis, We found that pTERTm carriers in cancer patients are older than noncarriers (Mean difference (MD) = 5.24; 95% confidence interval [CI], 2.00 to 8.48), male patients were more likely to harbour pTERTm (odds Ratios (OR) = 1.38; 95% CI, 1.22 to 1.58), and that pTERTm had a significant association with distant metastasis , a higher tumour grade in patients with glioma  and a higher tumour stage in other types of cancer . pTERTm was also significantly associated with a greater risk of death .pTERTm are a moderately prevalent genetic event in NSCLC. The current meta-analysis indicates that pTERTm is associated with patient age, gender and distant metastasis. It may serves as an adverse prognostic factor in individuals with cancers. The telomerase reverse transcriptase (TERT) gene encodes a highly specific reverse transcriptase that adds repeats to the 3′ end of chromosomes . The incThe relative characteristics and prognostic effects of TERT promoter mutation (pTERTm) on carriers and noncarriers with cancer are unclear. Statistical difference in gender distribution between pTERTm carriers and noncarriers was found in some studies that male cancer patients are more likely to harbour pTERTm ,10,11. RThe prevalence and association of pTERTms with non-small-cell-lung-cancer (NSCLC) patients have been studied but showed different results. Ma and colleagues found a proportion of 2.67% NSCLC patients in their cohort had pTERTm , whereasWe obtained 103 liquid nitrogen–stored tissue samples of 103 NSCLC patients with pathologic confirmation who were admitted to the First affiliated Hospital of Zhejiang University between 2013 and 2014. Sufficient high-quality tumour samples were taken at the time of surgical resection by well-trained physicians with written informed consent from each patient. Each sample was placed in liquid nitrogen immediately after resected and stored in -80°C refrigerator. Patient clinical data were collected and their information was anonymized and de-identified prior to this analysis. This cohort study was conducted under the approval of the Ethics Committees of the First affiliated Hospital of Zhejiang UniversityDNA extraction and polymerase chain reaction amplification for sequencing of the TERT promoter were performed in all cases by standard protocols. The genomic DNA of tumour tissue was extracted with a QiAamp DNA Mini Kit  and purified with an EZNA MicroElute DNA Clean-Up kit (OMEGA). Polymerase chain reaction (PCR) amplification of the TERT promoter region covering the mutations (from –27 to –286) was performed using primers: 5′ CCC ACG TGC GCA GCA GGA C3′ (forward) and 5′ CTC CCA GTG GAT TCG CGG GC3′ (reverse), With 3 minutes at 95°C; 35 cycles at 95°C 15 seconds, 63°C 15 seconds, 72°C 1 minute, followed by a final step at 72°C for 5 minutes. After gel electrophoresis to confirm the quality of the PCR products, sequencing PCR was performed using a Big Dye terminator version 3.1 cycle sequencing ready reaction kit (Applied Biosystems), and DNA sequence was analysed on an ABI PRISM 3730 automated genetic analyser (Applied Biosystems), All samples were checked in forward and reverse directions.Statistical analyses were carried out using the SPSS16.0 software package. Associations between pTERTm and the patients’ categorical variables were analysed with a chi-square test, Continuous data were summarised as the mean ± SD and analysed with the Mann-Whitney Wilcoxon test. Values of p less than 0.05 were considered significant.We searched PubMed and Web of Science for articles published before March 2015, using the systemic literature search terms “telomerase reverse transcriptase”, “promoter”, and “mutation”. The reference lists of the articles retrieved were further screened for other potential studies. We made every attempt to obtain the necessary information from the first and corresponding authors by e-mail if insufficient data were reported in the article .All of the studies included in this meta-analysis met the following criteria: (a) articles about the pTERTm and human cancer that were published in English. (b) availability of detailed genotype data or frequencies that could be calculated from the article text; (c) sufficient data to calculate an odds ratio (OR) or hazard ratio  with a 95% confidence interval (CI); (d) if survival data is not available for calculating HR, survival curves for pTERTm carriers and noncarriers is necessary. The exclusion criteria were: (a) published as an abstract, case report, comment letter, review or editorial; (b) non-human studies; (c) duplicate studies, in which case the latest or largest study were included.Two reviewers independently assessed all of the potentially relevant studies and reached a consensus on all of the items. Any disagreements were reconciled by discussion and consensus. The following data were collected from each study: first author, year of publication, type of cancer, population, sequencing method and the number of carriers and noncarriers.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp. This evaluation system focuses on three aspects of a study . The quality of the study was denoted by a numerical score from 0 to 9, with 9 representing the highest quality. Quality assessment was conducted by two independent reviewers. The original papers were scanned when disagreements occurred. Unsettled disagreements were referred to a third researcher for a final decision.The quality of the studies included was evaluated according to the Newcastle-Ottawa scale (NOS) quality assessment, which is available at The meta-analyses, subgroup analyses and sensitivity analyses were performed with Review Manager (revman) version 5.1 software. The meta-regression, Begg’s and Egger’s test were performed with STATA software .Heterogeneity > 0.05, or random effect model (DerSimonian and Laird method) [Heterogeneity < 0.05. Such as the assessment of association between pTERTm and gender , lymphatic metastasis (positive vs. negative), distant metastasis (positive vs. negative), tumour stage (III/IV vs. I/II), and Glioma WHO grade (III/IV vs. I/II). The dependent variables in these studies are the frequencies of event versus non-events. The significance of the combined OR was determined with a Z test, in which p < 0.05 was considered statistically significant. For continuous outcomes, the mean difference (MD) was calculated based on the mean and standard deviation given in the included studies. So the association between pTERTm and patient age at diagnosis was evaluated by mean age difference (carriers vs. noncarriers) combined with the corresponding 95% CIs. Pooled HR with a 95% confidence interval was calculated for the association between 5-year overall survival and pTERTm status (carriers vs. noncarriers). HR < 1 means that the prognosis of patients of pTERTm carrier is worse than non-carriers, while HR > 1 means the opposite. If a direct report of survival were not available, then the survival data read from Kaplan-Meier curves were read by Engauge Digitizer version 4.1 (http://digitizer.sourceforge.net/). Population data sets were categorized as Asian and non-Asian. Stratified analyses were performed by cancer type . The evaluation of the meta-analysis results included an examination of the heterogeneity, an analysis of the sensitivity, meta-regression and an examination for publication bias.For dichotomous outcomes, Odds Ratio (OR) with 95% confidence intervals was calculated by using a fixed effect model (Mantel-Haenszel method)  for PHet method)  for PHet2 was calculated to quantitatively estimate the heterogeneity, with I2 < 25%, I2 = 25–75% and I2 > 75% representing low, moderate and high heterogeneity, respectively. Subgroup and meta-regression were conducted to delineate the major sources of heterogeneity. Sensitivity analyses were performed to assess the stability of the results and to identify the individual potential influences on the OR or HR. Funnel plots and Egger’s test were used for the diagnosis of potential publication bias, An asymmetric plot suggests a possible publication bias and the P value of Egger’s test being considered representative of significant publication bias if it was less than 0.05.The heterogeneity between studies was evaluated using a chi-square–based Q test and a p value of less than 0.05 was considered statistically significant. The Higgins IThe study included 103 surgical specimens from patients with NSCLC. The results of the cohort study are shown in The detailed selection process is demonstrated in The overall results show that pTERTm carriers were older than noncarriers  from a random model. Stratification analysis decreased heterogeneity and identified increased MD in subgroup of glioma and lung cancer, whereas melanoma displayed a reversed pattern . No significant difference was found in other cancers. , S1 FigWe also found that male cancer patients were more likely to harbour pTERTm . But non-significant risk was found in glioma, lung cancer and renal cell carcinoma , S2 Fig.2 > 75). Sensitivity analysis with one study omitted each time showed that the significance of the result was not affected by any single study  Tables  and 5. SWe preformed meta-regression analyses by covariates including population, sample size, age, treatment, HR estimation and NOS score. No significant alteration was found in the HR by these covariates, and the results showed that the differences between the subgroups did not reach statistical significance . No evidBegg’s funnel plot and Egger’s test were both performed to evaluate the publication bias of the studies. The shapes of the funnel plots did not show any evidence of an obvious asymmetry in any comparison model. As shown in The maintenance of telomere length is of ultimate importance to normal self-renewing stem cells and cancer cells for preventing senescence induction. It has been suggested that tumour cells rely on epigenetic mechanisms or alterations that maintain telomerase activity to retain their immortality ,50,51. TThe prevalence of pTERTm in small-cell lung cancer (SCLC) and NSCLC have been investigated. Zheng et al  failed tIn the current meta-analysis, a borderline significant association between pTERTm and relevant clinical data was observed in overall analysis except for lymphatic analysis. The obvious between-study heterogeneity in each analysis decreased markedly in stratification analyses by tumour types, suggesting that different tumour types might be a potential source of heterogeneity. Interestingly, we observed a significant association of pTERTm with a higher age at diagnosis in patients with glioma and thyroid cancer, whereas patients with melanoma displayed an opposite pattern. This is probably because genetic factors and environmental factors contribute equally to the development of melanoma. Recent studies suggested that melanoma is found more frequently in skin with intermittent sun-exposure than in skin that is not exposed or chronically exposed ,54.In addition, we found that thyroid cancer patients with pTERTm have a higher risk of distant metastasis that is four times greater than that of patients without pTERTm , in line with the study done by Gandofi et al. They found that pTERTm are strongly associated with tumour progression and development of distant metastasis in papillary thyroid cancer . SimilarThe 5-year overall survival data from 25 studies indicated that patients with pTERTm had a 70% greater risk of death than those without pTERTm. Since pTERTm results in the creation of binding sites for ETS/TCF transcription factors, which are downstream targets of RAS-RAF-MAPK pathways. pTERTm are suggested to have synergistic effects to promoter tumour cell proliferation with activating BRAF or NRAS mutations, which have been proposed to be driver mutations in the development of cutaneous melanocytic neoplasms. It is likely that these mutations turn the pTERTm into a target of ETS-domain transcription factors. Thus additional studies could further investigate whether pTERTm are of therapeutic significance, either in terms of influencing the efficacy of established therapies  or whether they might even prove to be directly valuable to therapeutic targets,58,59. TThe funnel plots and Egger’s test did not identify any publication bias. However, some limitations should be addressed in the interpretation of the results of our cohort study and meta-analysis. First, the sample size of our cohort study was relatively small. Well-designed population-based studies with large sample sizes and detailed exposure information are needed to further confirm our findings. Second, subgroup meta-analysis stratified by cancer type, such as hepatocellular carcinoma, bladder cancer and laryngeal cancer, might contain insufficient data to enforce statistical power to check for an association, despite our efforts to contact the authors for data. We were unable to include more articles because the authors of a few studies with incomplete data failed to reply to our requests. Hence, more individual study would be required to draw a more precise conclusionIn conclusion, we found that pTERTm is present in a small fraction of NSCLCs and are significantly associated with older patients. The meta-analyses suggested that pTERTm carriers were older than noncarriers in glioma, thyroid cancer and lung cancer, with melanoma demonstrate a reserved pattern. Male cancer patients exhibited a significantly elevated risk of having pTERTm in thyroid cancer, melanoma and hepatocellular carcinoma. Apart from other cancers, we also identified thyroid cancer patients with hTERTm are more likely to have distant metastasis and higher tumour stages. In addition, pTERTm carriers had a higher risk of death in our prognosis analysis in giloma, thyroid cancer, gynecologic cancer and “other cancers”. All in all, the detection of pTERTm appears to be a promising prognostic indicator in patients with cancer and may have potential as a biomarker for treatment stratification. More well-designed prospective studies are needed to validate our findings.S1 PRISMA Checklist(DOC)Click here for additional data file.S1 Fig(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S3 Fig(TIF)Click here for additional data file.S4 Fig(TIF)Click here for additional data file.S5 Fig(TIF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(DOCX)Click here for additional data file.S4 Table(DOCX)Click here for additional data file.S5 Table(DOCX)Click here for additional data file.