Studies investigating the contribution of Cytochrome P4502E1 (CYP2E1) polymorphisms to the etiology of urinary cancer draw inconsistent conclusions. Thus, we performed a meta-analysis to evaluate the association between CYP2E1 Rsa I/Pst I and Dra I polymorphisms and urinary cancer susceptibility.Meta-analysis based on the eligible case-control studies that assess the association of CYP2E1 Rsa I/Pst I and Dra I polymorphisms with urinary cancer was conducted. Subgroup analyses based on ethnicity and cancer type were also carried out. Odds ratios (OR) and 95% confidence intervals (95% CI) were calculated to evaluate the strength of the associations between the two polymorphisms. Funnel plot and Begg’s test were used for publication bias diagnosis.We found decreased urinary cancer risk among subjects carrying CYP2E1 RsaI/PstI c1c2 + c2c2 genotype and c2 allele , with 3,301 cases and 3,786 controls from 14 studies. We also observed a significant difference in c1c2 + c2c2 vs. c1c1 and c2 vs. c1 among Asians . However, the meta-analysis based on 5 eligible studies showed no significant association between CYP2E1 Dra I polymorphism and urinary cancer susceptibility in either dominant model or the allele model.Our meta-analysis concluded that CYP2E1 Rsa I/Pst I polymorphism correlates with urinary cancers risk in Asian population; while CYP2E1 Dra I polymorphism might be not significantly associated with the urinary cancer risks. Large and well-designed studies are needed to confirm these results. Cancer is a leading cause of death worldwide, which accounted for 7.6 million deaths in 2008 . ProstatRecently, a number of gene polymorphisms which were associated with urinary cancer risk have been found, and some polymorphisms located in the genes associated with carcinogen metabolism courses . Human cCytochrome P4502E1 (CYP2E1) is belonged to the member of the cytochrome P450 superfamily. It is a phase I enzyme that could metabolically activate many kinds of carcinogens , 9. N-nihttp://www.ncbi.nlm.nih.gov/SNP). Though, only a few polymorphisms which might alter the enzymatic activity of CYP2E1 have drawn interest [G > C is associated with Pst I restriction enzyme site while rs2031920 C > T accounts for Rsa I restriction enzyme site. The two polymorphisms were in complete linkage disequilibrium, which results in the CYP2E1*5 haplotype and three different genotypes: homozygous of normal alleles , heterozygous (c1c2) and homozygous after nucleotide replacement [CC, CD and DD [CYP2E1 gene is located on chromosome10q26.3 and is consist of nine exons and eight introns. So far, over one hundred single nucleotide polymorphisms (SNPs) have been reported . Several /Pst I+). AnotherD and DD . This poD and DD  and was D and DD . TherefoRecently, the associations between CYP2E1 gene and urinary cancer risk were investigated. However, the results from epidemiological studies were inconsistent and controversial –31. The We carried out a comprehensive search in the databases PubMed, Elsevier, SpringerLink and CNKI  without a language limitation, covering all the papers published up to Oct 2016. The search strategy to identify all possible studies involved used combinations of the following key words: (cytochrome P4502E1 OR CYP2E1) and  and . A cited reference search of the retrieved papers was also conducted, and further publications were also identified by retrieving the bibliographies of the retrieved papers.Data from studies were included in this meta-analysis only if the study met the following criteria: (1) the study should concern the association of CYP2E1 RsaI/PstI or Dra I polymorphisms with urinary cancer risk; (2) only the observational (case-control or cohort) studies in accordance with Hardy-Weinberg equilibrium (HWE) were considered; (3) the paper must indicate the sample size, odds ratios (ORs) and their 95% confidence intervals (CI) as well as the genetic distribution or the information that can help infer the results. Accordingly, papers that could not offer the source of cases and controls or other essential information were excluded; reviews, editorial and comments were also excluded. After rigorous searching, we reviewed all the papers based on the above criteria for further analysis.Data were carefully extracted from all the eligible publications by two investigators independently according to the inclusion criteria. For each study, the following information was extracted from the study: name of first author, year of publication, region, ethnicity, gender, cancer types, matching criteria, the numbers of cases and controls with the three genotypes and genotyping methods. When the essential information was not provided in articles, every effort was made to contact the authors. In the case of conflicting evaluations, disagreements of included studies were resolved by discussion. When a consensus cannot be reached, another author was to be consulted to resolve the dispute, and then a final decision was made based on a majority of votes.P value > 0.05 were considered to be in accordance with HWE. We then used Chi-square test to calculate the genotype and allele difference of CYP2E1 Rsa I/Pst I and Dra I polymorphisms in Caucasian and Asian populations. For CYP2E1 Rsa I/Pst I and Dra I polymorphisms, we only assessed the risk in the dominant model and the allele model due to few frequencies of mutated genotypes in subjects. Afterwards, we performed subgroup meta-analysis according to the status of ethnicity and cancer type. The pooled OR with 95% CI was calculated to evaluate the associations of CYP2E1 Rsa I/Pst I and Dra I polymorphisms with urinary cancer risk, using subjects with the homozygous common allele as the reference group. The significance of the overall OR was determined by the Z-test.HWE for CYP2E1 Rsa I/Pst I and Dra I polymorphisms of control groups were extracted from the original publications. In case of studies without reporting HWE status, the distributions of genotypes were tested for HWE using the Chi-square test and results with Q test [2 test [P < 0.10 or I2 value ≥ 50% was considered to represent significant heterogeneity. The random effects model (the DerSimonian-Laird method), which yields wider confidence intervals, would be adopted to calculate the overall OR value if the test of heterogeneity was significant [Cochran’s Chi-square based Q test  and I2 t [2 test  were adonificant ; otherwinificant . Publicanificant  with STAP value less than 0.05 was considered to be statistically significant.All of the statistical analyses were performed with RevMan  and STATA (Version 12.1). All the tests were two-sided and There were a total of 44 studies preliminarily identified by searching the PubMed, Elsevier, Springer Link and MEDLINE databases. After screening titles and abstracts, 29 studies were identified to be relevant. After full text searching, 15 studies were excluded. According to the eligible criteria, three studies were discarded for insufficient data , two studies were discarded as being case-only studies and ten studies were discarded as being review articles. Finally, 14 studies were identified for the CYP2E1 Rsa I/Pst I polymorphism, including a total of 3,301 cases and 3,786 controls, and for the DraI polymorphism 5 studies were identified covering a total of 1,168 cases and 1,275 controls. The flow diagram of selection strategy was shown in Figure P value < 0.001). As for CYP2E1 Dra I polymorphism, the frequency distributions of the genotypes and alleles were also statistically significant between the Caucasian and Asian groups .Firstly, we evaluated the genotype and allele frequency distributions of CYP2E1 Rsa I/Pst I and Dra I polymorphisms between Caucasian and Asian control populations Table . As for P < 0.001) as well as the allele model . Considering the ethnic discrepancy in the genotype/allele frequency of the polymorphism, we studied the impact of CYP2E1 RsaI/PstI in Asian and Caucasian population. A significant difference in c1c2+c2c2 vs. c1c1 and c2 vs. c1 among Asians was observed, with the summarized ORs being equal to 0.682 (95% CI = 0.601–0.784) and 0.751 (95% CI = 0.663–0.851) respectively. However, we failed to observe any difference among Caucasians in either dominant mode  or allele model . Summarized ORs for CYP2E1 RsaI/PstI stratified by cancer type were also evaluated. We observed subjects with c1c2/c2c2 genotype had decreased risk than those with c1c1 genotype for developing prostate, urothelial and bladder cancer, with the summarized ORs being equal to 0.609 , 0.773  and 0.810  respectively. Consistently, similar results were observed with the allele model in the subgroup analysis considering cancer type. Subjects with the c2 allele had decreased susceptibility than those with the c1 allele for prostate, urothelial and bladder cancer, with the summarized ORs being 0.710 , 0.794  and 0.832 , respectively.The main results about CYP2E1 RsaI/PstI polymorphism were summarized in Table P = 0.202) or the allele model  was showed. Further subgroup analysis considering ethnicity and cancer type was conducted. We demonstrated that prostate cancer patients carrying CD/CC genotype had 0.66-fold risk than patients with DD genotype . Other subgroup summarized ORs were found no significant difference  (Table The principle results of CYP2E1 Dra I polymorphism were shown in Table 5) Table .P < 0.10 or I2 ≥ 50%), and all the P values and I2 values were listed in Tables In the meta-analysis, no significant heterogeneity was found in any of the comparisons (P > 0.05).Publication bias was firstly examined by using funnel plot analysis Figure . As a coIt is widely accepted that genetic susceptibility plays a key role in the development of urinary cancer, though the underlying mechanism of urinary cancer has not been clearly illuminated , 5. A vaAccording to previous reports, six restriction fragment length polymorphisms (RFLPs) were resides in CYP2E1 gene . AmongstP < 0.001), comparing with the subjects carrying wide-type homozygous c1c1 genotype. In the subgroup analyses sorted by cancer category, significant associations were found in prostate cancer, urothelial cancer as well as bladder cancer group. Furthermore, ethnicity may influence CYP2E1 activity through gene–gene interactions since it is an important biological factor [As for CYP2E1 Rsa I/Pst I polymorphism, the results have demonstrated the mutated allele c2 may have different effects in different cancer type –46. In rP = 0.202; C vs. D, OR = 0.978, 95% CI = 0.847–1.117, P = 0.748). Further subgroup analysis sorted by the ethnicity and cancer category indicated that patients carrying CD/CC genotype had 0.658-fold risk compared with patients with DD genotype to develop into prostate cancer . Nevertheless, no significant association was found for Asians, Caucasians and other cancer types in neither genetic model. Since sample size is a key factor for identifying risk factors, we may undervalue the association between Dra I polymorphism and urinary cancer risk considering the limited quantity of eligible studies.As for CYP2E1 Dra I polymorphism, we identified five studies that had examined the relationship between CYP2E1 Dra I polymorphism and urinary cancer. The pooled result failed to identified any significant associations between CYP2E1 Dra I polymorphism and the urinary cancer risk  assay demonstrated the enhancement of expression by c2c2, which was about 10 times than that by its c1c1 counterpart. The DNase I sensitivities and protection profiles of the two genotypes were also different, with RsaI polymorphism affecting the transcriptional activation of CYP2E1 as well as the binding capacity of transcription factor HNF-1, while the PstI polymorphism having no impact . The stuThis is the first meta-analysis assessing the association between CYP2E1 Rsa I/Pst I as well as Dra I polymorphisms and the susceptibility of urinary cancers. Nevertheless, when interpreting the results of the meta-analysis, limitations must be taken into account. Firstly, only five studies with 1,168 cases and 1,275 controls reported the association between CYP2E1 Dra I polymorphism and urinary cancer risk. Therefore, significant associations might be underpowered considering the limited studies involved. More studies with larger sample size are necessary to get a more precise conclusion. Secondly, owing to the relatively low frequency of mutated alleles of CYP2E1 polymorphism in Caucasians, we failed to distinguish any significant association in the Caucasians subgroup . As a coIn the present meta-analysis, we provided preliminarily genetic evidence that CYP2E1 Rsa I/Pst I polymorphism was related with the susceptibility of urinary cancers in Asian population, though Dra I polymorphism failed to contribute to the susceptibility of urinary cancers except for prostate cancer. However, our findings have to be interpreted with caution due to several limitations. Studies involved large number of participants and more homogeneous cancer patients should be future conducted.


Previous observational studies have suggested a potential relationship between osteoarthritis (OA) and the risk of cardiovascular disease (CVD), with conflicting results. We aimed to provide a systematic and quantitative summary of the association between OA and the risk of CVD. We searched Medline and EMBASE to retrieve prospective and retrospective studies that reported risk estimates of the association between OA status and CVD risk. Pooled estimates were calculated by a random effects model. The search yielded 15 articles including a total of 358,944 participants, including 80,911 OA patients and 29,213 CVD patients. Overall, the risk of CVD was significantly increased by 24%  in patients with OA compared with the general population, with no significant publication bias. Furthermore, sensitivity analysis indicated that our results were robust and were not influenced by any one study. In conclusion, this meta-analysis provides strong evidence that OA is a significant risk factor for CVD. Cardiovascular disease (CVD), such as ischemic heart disease (IHD), congestive heart failure (CHF), transient ischemic attacks (TIA), and stroke, is a leading cause of morbidity and mortality in the general population worldwide. According to the World Health Organization, 17.5 million people die each year from CVD, constituting approximately 30% of all deaths worldwideOsteoarthritis (OA) is also a major cause of morbidity and healthcare expenditures and affects approximately 15% of the populationet al. found a linear association between the severity of hand OA and atherosclerosis in the AGES Reykjavik studyet al. concluded that symptomatic hand OA but not radiographic hand OA was associated with an increased risk of coronary heart disease events using the data from the Framingham Heart StudyRecent epidemiological studies have suggested a potential relationship between OA and CVD, with conflicting results. Whereas Jonsson Overall, 3,075 references were initially identified. After the initial screening of titles and abstracts, a total of 3,028 articles were excluded, leaving 47 articles for retrieval. Full text assessment of these articles resulted in 15 eligible articles that met our inclusion criteria45678910111213141516174912145678101113151757104689111215456891012131415161748137Features of exposure varied across studies. Although 12 studies ascertained OA by either radiographic results or medical records456789111213171014468116111767The method of outcome ascertainment varied across articles. Although the majority of articles ascertained CVD by medical records, three articles used questionnaires to ascertain CVD4146514781011134915Although the included articles were quite heterogeneous, the inter-reviewer reliability for data extraction was almost perfect (kappa ranged 0.99–1.00). All studies were rated as medium to high quality.Fifteen articles provided 19 risk estimates of the association between OA and CVD risk. Of these 19 estimates, 11 reported that OA was associated with a significantly increased risk of CVD, five that OA was associated with a non-significantly increased CVD risk, and three that OA was associated with a non-significantly decreased CVD risk. No study included in this systematic review reported that OA was associated with a significantly decreased risk of CVD.For the non-parametric sign test, we considered whether studies reported an increased or decreased risk of CVD associated with OA, regardless of significance level or magnitude of effect. Overall, 16 risk estimates found that OA was associated with an increased risk of CVD, and three found that OA was associated with a decreased risk of CVD. The sign test (P = 0.004) rejected the null hypothesis of equal CVD risk in patients with and without OA.I2 = 85.1%. Overall, the risk of CVD was significantly increased by 24%  in patients with OA compared with the general population. No significant publication bias was observed according to the Egger test (P = 0.659) or funnel plot  had a pooled RR of 1.30  and the retrospective studies (seven estimates) had a pooled RR of 1.15 .Second, we excluded studies that ascertained OA or CVD status by questionnaire. The remaining studies had a pooled RR of 1.26 . Finally, we excluded individual study estimates one at a time to examine the influence of each study on the overall RR. We found that the omission of any one study did not appreciably change the pooled RR .When subgroups were stratified by study design, we found a significant effect of prospective studies  and a non-significant effect of retrospective studies . The stratified analysis according to OA site included 3 subgroups, hand , knee  and hip . When studies were stratified according to confirmation methods for OA, we found a RR of 1.47  for clinical OA and a RR of 1.23  for radiographic OA. When studies were stratified according to different CVD types, we found significant results for the IHD , CHF  and cardiovascular death  subgroups, whereas a non-significant result was found for the stroke subgroup . Finally, we considered whether subgroup analysis reported a higher or lower CVD risk associated with OA, regardless of significance level. The results indicated that all subgroup estimates found that OA was associated with an increased risk of CVD. The results of the subgroup analyses are summarized in The impact of age and follow up time on the pooled result was explored by meta-regression analysis. As illustrated in This is the most comprehensive systematic review and meta-analysis of published observational studies assessing the relationship between OA and CVD risk. The results of our meta-analysis demonstrate that OA is associated with a significantly increased risk of CVD. Furthermore, this association was robust across sensitivity analyses that accounted for the influence of each individual study.The underlying mechanisms behind the observed association between OA and CVD risk remain unclear, but several factors may account for this relationship. First, the two diseases have some shared risk factors. Epidemiological studies have provided evidence for an association between OA and most of the traditional cardiovascular risk factors, including hypertension1926et al.et al.Potential limitations of this meta-analysis arise from the unavailability of individual participant data from the included studies. For instance, Jonsson The strengths of this study include the comprehensive and systematic review of the literature. Study selection, data extraction, and quality assessment were performed independently by two authors according to predesigned criteria to minimize bias and transcription errors. We included both prospective and retrospective studies that used large sample sizes, which increased the statistical power to detect potential associations. Compared to the previous meta-analysis on this topicBecause OA is a very common health condition, an association between OA and CVD would be important from a public health perspective. In the general population, middle-aged people may consider screening for OA status as well as traditional cardiovascular risk factors to enable early intervention to reduce future CVD events. Patients with OA should pay more attenuation to their CVD risk. Among clinicians, cardiovascular risk must be taken into account when prescribing any non-steroidal anti-inflammatory drug for OA patients.In conclusion, this meta-analysis provides strong evidence that OA is a significant risk factor for CVD. Given the high prevalence and incidence of OA and CVD in the general population, the observed relationship between OA and CVD has clinical and public health importance.We carried out a meta-analysis of studies that evaluated the association between osteoarthritis and cardiovascular diseases in adults. We followed the quality of reporting of meta-analysis guidelines (the PRISMA statement) for performing and reporting the present meta-analysis  and S229The eligibility of studies was assessed through a three-step process. First, two independent reviewers performed an initial screening of all titles and abstracts according to the following criteria: (1) original research articles were included, and other types of articles, including reviews, editors, commentaries and meta-analyses, were excluded; (2) population-based association studies reporting the relationship between OA and CVD risk were included, and articles focused on other exposure or outcomes were excluded. The full texts of all potentially relevant articles were then reviewed, and studies were included if they met the following criteria: (1) predefined diagnosis criteria for both OA and CVDs; (2) reported risk ratio (RR), hazards ratio (HR), or odds ratio (OR) estimates and 95% CIs describing the relationship between OA and risk of CVDs; (3) inclusion of a non-exposed group in prospective studies or a control group in retrospective studies. Finally, discrepancies were resolved by consensus or consultation with a third reviewer.We designed a standardized data collection form to extract information. Two reviewers independently performed the extraction of data. We adopted OR for retrospective studies and HR and RR for prospective studies as a measure of the association. The following characteristics were recorded in the data abstraction form: study name, authors, year of publication, residential region, type of study, source of the study population, OA and CVD definition, sample size, and adjusted confounding factors.After the data extraction procedure, the two forms from the two reviewers were compared in a point-by-point manner, and the degree of agreement between the reviewers was assessed using kappa statistics. Any discrepancies were resolved by consensus with a third investigator while referring to the original article.Finally, we assessed study quality according to the Newcastle–Ottawa quality assessment scale, which was recommended by the Cochrane guidelines. Details of the scoring system and the quality score of each study are listed in For each study, we extracted the estimate and 95% confidence interval for the association result. The ORs in retrospective studies were converted to RRs for meta-analysis (RR = OR/([1 − pRef] + [pRef * OR]), where pRef is the prevalence of CVD in the control groupWe calculated a pooled RR estimate across all studies using a random-effects model that assumes that individual studies estimate different association effects. The random-effects model was adopted because it is probably the most conservative analysis to account for variance within and between studies.I2 statistic, calculated from the Q statisticI2 > 75% was considered a high level of heterogeneity.The between-study heterogeneity was assessed by means of the Publication bias was assessed by visually examining the asymmetry of a funnel plot in which the log estimates were plotted against their standard errors. Furthermore, we also employed an Egger regression test in our analysis to calculate two-tailed P values for quantifying publication biasTo explore the influence of potential sources of heterogeneity on the pooled estimate, we carried out subgroup analyses by important study characteristics, including study design (prospective or retrospective), OA sites , OA types , and outcomes . The impact of mean age and follow up time on the pooled result was explored by meta-regression. Analyses were performed with Stata Version 12.0 .How to cite this article: Wang, H. et al. Osteoarthritis and the risk of cardiovascular disease: a meta-analysis of observational studies. Sci. Rep.6, 39672; doi: 10.1038/srep39672 (2016).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


Genetic polymorphisms, including single nucleotide polymorphisms (SNP) and nucleotide repeat expansions, can occur in regions that transcribe non-coding RNAs (ncRNA), such as, but not limited to, micro RNA and long non-coding RNA. An association between genetic polymorphisms of ncRNA and increasing head and neck cancer (HNC) risk has been identified by several studies. Therefore, the aim of this systematic review is to consolidate existing findings to clarify this association. Four electronic databases, such as MEDLINE, EMBASE, Chinese BioMedical Literature Database, and China National Knowledge Infrastructure, were utilised. Inclusion of studies and data extraction were accomplished in duplicate. A total of 42 eligible studies were included, involving 28,527 cases and 37,151 controls. Meta-analysis, sensitivity analysis and publication bias detection were performed. Among the eligible studies, 102 SNPs were investigated, and 21 of them were considered eligible for meta-analysis. Our analysis revealed that HOTAIR rs920778, uc003opf.1 rs11752942, and miR-196a2 rs11614913 were related to HNC susceptibility, while let-7 rs10877887, miR-124-1rs531564, and miR-608 rs4919510 were considered as protective factors. In conclusion, our results showed the extreme importance of an up-to-date comprehensive meta-analysis encompassing the most recent findings to obtain a relevant and reliable framework to understand the relationship between ncRNA SNPs and HNC susceptibility. Head and neck cancer (HNC) is a broad term encompassing epithelial malignancies located in the oral cavity, pharynx, larynx, cervical oesophagus, paranasal sinuses, nasal cavity, salivary glands and thyroid gland –3. The aPervasive transcription of human genomes has been demonstrated with the development of whole genome technologies. Interestingly, only 2% of transcripts encode proteins, while most transcripts are ncRNAs –11. ncRNMiRNAs are small ncRNAs of 21–23 nucleotides, which use partial base-pairing to recognize target messenger RNAs, thus repressing their expression . Hence, Despite physical differences, human populations are 99% genetically identical, which means that human diversity, including cancer susceptibility, results from the remaining 1% of genetic polymorphisms –21. SmalThe process of study identification was showed in After the removal of duplicates, 193 records were identified through electronic database searching and manual searching. One hundred and thirty-one records were excluded after the screening of the abstract, while the full texts of the remaining 62 records were analysed to confirm their eligibility. Among these 62, 20 were excluded, such as three studies in which primary head and neck cancer was actually not diagnosed, seven studies that did not investigate ncRNA polymorphisms, six studies that did not analyse the association between genetic polymorphism and cancer risk, and four studies that did not report sufficient data for statistical analysis. Finally, 42 studies –69 were Among the 42 included studies, six studies , 62, 64 P > 0.05). Except all the three cohorts of miR-219-1 rs213210 and two of the four cohorts of miR-26a-1: rs7372209, the rest cohorts were of different SNPs  of included studies was listed in Meta-analysis results are listed in A total of 19 SNPs found into miRNAs gene loci were included into meta-analysis. The overall pooled result showed a significant association between miRNA SNP and increased HNC risk only in 1 SNP, such as miR-196a2 rs11614913 (C/T), in co-dominant model , which was supported by 13 comparisons (7107 cases and 7233 controls). Furthermore, we found that several SNPs acted as protective factors according to the overall pooled result in certain genetic models, such as let-7 rs10877887 ,miR-124-1 rs531564, and miR-608 rs4919510, which were supported by two comparisons (1002 cases and 1293 controls), two comparisons (1738 cases and 1961 controls), and four comparisons (3191 cases and 3899 controls), respectively.n = 2). According to the methods described in methods part, subgroup analysis was performed based on ethnicity and cancer type. The results of subgroup analysis were listed in Statistical heterogeneity was found in 11 SNPs. Eight SNPs underwent subgroup analysis to detect the source of heterogeneity, while miR-218-2 rs11134527 (G/A), and miR-27a rs895819 (T/C), miR-449b rs10061133 (A/G) were not subjected to subgroup analysis because the number of included studies was too small(MiR-146a rs2910164 (G/C) was found to be related to NPC susceptibility , while only two comparisons (393 cases and 373 controls) were included in this subgroup. We also found that miR-196a2 rs11614913T allele could significantly increase HNC risk in Asian populations  supported by 10 comparisons (5189 cases and 5213 controls). SNP miR-196a2 rs11614913 was also found to be related to OSCC susceptibility by pooled result of four comparisons (1561 cases and 1304 controls). Protective factors were also evaluated in specific ethnic backgrounds or cancer types. SNP miR-608 rs4919510 (G/C) showed protective effect on NPC susceptibility , while only two comparisons (1590 cases and 1979 controls) were included in this subgroup. In addition, miR-499a rs3746444 (T/C) was found to be a protective factor in Caucasian populations , and contributed to ESCC , as supported by the pooled result of two comparisons (669 cases and 688 controls).Interestingly, the overall pooled result for miR-34b/c rs4938723 (T/C) differed from the subgroup analysis result. The overall pooled result of four comparisons (2739 cases and 3327 controls) indicated that SNP miR-34b/c rs4938723 (T/C) was associated with HNC susceptibility , while subgroup analysis showed its protective effect against ESCC , as supported by pooled result of two comparisons (1738 cases and 1961 controls).In addition, omitting the cohorts whose control group was not in agreement with HWE did not change the result of meta-analysis (data not shown). Besides, omitting the cohorts that were considered as high risk of bias did not changed the results of meta-analyses (data not shown). The influence of cohort that was considered as medium risk of bias on the pooled result would be discussed in conclusion below.No significant change was observed when omitting individual studies in turn, suggesting that the results were reliable. Publication bias of the two SNPs, such as miR-146a rs2910164 and miR-196a2 rs11614913 was detected by funnel plots . AsymmetIdentifying genetic markers of cancer susceptibility and gene-environment interactions could contribute to the reduction of cancer mortality through early diagnosis and personalized therapy . Genome-According to a recent study by the ENCODE Project Consortium, approximately 9640 lncRNA gene loci are present in the human genome, while the total number continues to grow –73. The A large number of studies have proved that HOTAIR can act as an oncogene –76. HOTABiogenesis and maturation of miRNAs is well depicted by several studies over the past decades –83. Brieet al. [. Hence, to the best of our knowledge, the present systematic review represents the most comprehensive meta-analysis since it incorporated the most recent studies when compared to other reviews associated to miRNA SNPs and HNC susceptibility relationship.Large numbers of systematic reviews have been performed to investigate the association between miRNAs SNPs and cancer susceptibility, including breast cancer, colorectal cancer, gastric cancer, liver cancer, lung cancer –90. In cet al. . Hence, The SNP rs2910164 in miR-146a is located in the stem region of the precursor opposite to the mature miR-146 sequence, therefore a change from G allele to C allele could affect the stem structure of the miR-146a precursor, thus decreasing mature miRNA expression and suppressing oncogenesis , 92. TheMiR-196a2 is composed of two different mature miRNAs, such as miR-196a-5P and miR-196a-3P, both processed from the same stem-loop . Rs11614SNPs resident in miRNAs gene loci could not only increase cancer risk, but also play protective roles in cancer development. In the present work, we found four miRNA related SNPs, such as miR-499a rs3746444 (T/C), miR-608 rs4919510 (G/C), let-7 rs10877887 (T/C), and miR-124-1 rs531564 (C/G), significantly associated with decreased HNC risk after meta-analysis.in vitro [The correlation between SNP rs3746444 in miR-499a and HNC susceptibility was broadly investigated, thus a total of 9 studies were included into our meta-analysis. Although the overall pooled results were not statistically significant, subgroup analysis demonstrated that this SNP was associated with decreased HNC risk, as well as ESCC risk, in Caucasian populations. The rs4919510 is located in the miR-608 mature sequence, thus the mutant C-allelic mir-608 exerts a significantly lower colony formation Let-7 is functionally considered as a tumour suppressor gene due to its role of repressing RAS family , 95. SimOn the other hand, the overall pooled result demonstrated that miR-34b/c rs4938723 (T/C) was a risk factor for HNC in the co-dominant model, but showed a protective effect on ESCC development in the recessive model. The reason for this divergence is difficult to understand. On the basis of our results we could only postulate that the same miRNA SNP could exert different effects in different cancer types.Taken together, our present meta-analysis highlighted that systematic and comprehensive literature searching and inclusion, large sample size inclusion, consistency with HWE in almost all of the included cohorts, and reliable meta-analysis results confirmed by sensitive analysis and subgroup analysis, represent important and adequate elements to create a more comprehensive and reliable framework for understanding the potential roles of ncRNA polymorphisms in determining HNC susceptibility. However, there were still some limitations in this systematic review. Firstly, we included oesophageal cancer despite our intention to analyse HNC risk. Only cancers originating from the cervical oesophagus, which begins at the level of cricopharyngeus muscle and extends down to the sternal notch, could be anatomically considered as a HNC subtype. However, after full text screening, we did not find any study investigating ESCC related ncRNA SNPs related to each anatomical cancer location, such as cervical oesophagus, thoracic oesophagus, and abdominal oesophagus. After discussion, we decided to include these studies with the objective to provide a wider perspective for understanding HNC. Secondly, statistical heterogeneity was significant in some meta-analyses, which might be due to the difference between studies in cancer types and ethnic backgrounds. Thirdly, the present meta-analysis was based on unadjusted OR estimates. Lack of sufficient information made a more precise evaluation with ORs adjusted by other covariates including age, gender, smoking status, viral infections, difficult. Fourthly, since quite a few meta-analyses were performed, multiple comparison problem should be considered carefully. Multiple comparison problem may occur when several statistical interferences were considered simultaneously. In this systematic review, five genetic models were used, it may increase the risk of incorrect rejections of null hypotheses, i.e. false positive or type I error. If only one of the five genetic models showed positive result, such as miR-34b/c rs4938723, miR-196a2 rs11614913 and miR-608 rs4919510, the reliability of the result should be considered. Finally, the methodological quality assessed by NOS of several included studies was considered low, which indicated high risk of bias. Inadequate methodological design frequently occurred in the selection of controls. Controls should be frequency-matched in confounding factors, including age, gender and ethnic backgrounds with cases. Rather than just reporting no difference between case and control, the consistency of baseline should be ahead of the selection of controls, or adjust the OR for the confounding factors. Besides, the response rate of case and control was barely reported, and several studies were hospital-based, which could both lead to an increased risk of selection bias. Despite these limitations, our work revealed that an updated and comprehensive meta-analysis is of utmost importance for obtaining reliable results.The protocol of the present systematic review was approved by the institutional review board. Study selection, data extraction and quality assessment were performed by two authors (Pan W and Wu C) in duplicate, and discrepancies were resolved through consensus discussion.Studies that met the following inclusion criteria were included in the present systematic review: (1) cohort study or case-control study as study design; (2) patients with diagnosed HNC; (3) controls without any cancer history; (4) study that evaluated the association between ncRNAs polymorphisms and HNC susceptibility; and (5) collect an amount of sufficient data to calculate odds ratios (OR) and 95% confidence intervals (95% CI). Furthermore, patients risk suffering a secondary primary cancer, which might be the result of field cancerization. Based on this, malignancies located in oral cavity, pharynx, larynx, cervical oesophagus, paranasal sinuses, nasal cavity, salivary glands and thyroid gland, were preferred to be considered together. Thus, in the present systematic review, the malignancies mentioned above were regarded as HNC and included.The following electronic databases were employed: MEDLINE , EMBASE , Chinese BioMedical Literature Database , and China National Knowledge Infrastructure . The search strategy we used combined Medical Subject Headings (MeSH) with free text words. The MeSH terms used were “RNA, Untranslated”, “Polymorphism, Genetic” and “Head and Neck Neoplasms”. Relevant journals and reference lists of included studies were manually searched. The titles and abstracts of all studies were initially scanned to find any eligible study. The full text of the potential eligible studies was obtained to confirm their eligibility.http://www.ncbi.nlm.nih.gov/projects/SNP) as the original allele, and the variant alleles at the same genetic locus as exposure factor.For each included study, the following information was extracted: 1) first author name and publication year; 2) country of origin; 3) ethnicity, such as Asian, Caucasian and Black; 4) study design; 5) cancer type; 6) cases and controls number; 7) important information regarding the recruiting process; 8) candidate ncRNAs, with the ancestral allele of a certain genetic loci; and 9) genotype distribution of cases and controls. The data of each cohort were extracted separately and the cohorts were considered as different comparisons in further statistical analyses when more than one cohort was reported in one single study, including cohorts from different countries and cohorts from different regions of one country. Furthermore, since the minor allele frequency in control group varied in different ethnicity background, we decided to take the ancestral allele found in the dbSNP database  was chosen as the tool to assess the methodological quality of included studies. Studies rated 0–3 stars would be considered as high risk of bias, while 4–6 stars 7–9 stars and would be considered as medium and low risk of bias.NOS (available at: Hardy-Weinberg equilibrium (HWE) was used to assess genotype distribution of controls and a goodness-of-fit test (chi-square or Fisher exact test) was performed assess the departure from HWE.P < 0.05 was considered statistically significant. Subgroup analysis was also performed based on ethnicity background and cancer type.The meta-analysis was performed using RevMan version 5.3 from Cochrane Collaboration, and pooled ORs with 95% CIs were calculated separately in different genetic models to evaluate the association between ncRNA polymorphisms and HNC risk. Using SNP C>T as an example, OR and 95% CI were calculated in four genetic models, including allele contrast (C vs. T), dominant model (CC vs. CT + TT), recessive model (CC + CT vs. TT), and co-dominant model . The significance of the pooled OR was determined by the 2-tailed z-test and 2 statistics. When P > 0.05 and I2 < 50%, a fixed-effects model was applied; when p ≤ 0.05 or I2 ≥ 50%, a random-effects model was applied. To ensure the reliability of the results, sensitivity analysis was performed by omitting each study in turn. Publication bias was detected by the funnel plot when the number of included studies was more than 10, since limited number of included studies would affect the power of the tests to distinguish chance from real asymmetry. Harbord test was chosen as test for funnel plot asymmetry. The trim and fill method was applied to identify and correct for funnel plot asymmetry arising from publication bias. Sensitive analysis and detection of publication bias were conducted with STATA Version 14.0 software .Statistical heterogeneity between studies was analysed by Cochran’s Q statistic following a chi-square distribution and I


SMAD7) inhibits the transforming growth factor-β (TGF-β) signaling pathway, which regulates carcinogenesis and cancer progression. A number of studies have reported that SMAD7 polymorphisms  are associated with colorectal cancer (CRC) risk, but the results from these studies remain conflicting. To determine a more precise estimation of the relationship between SMAD7 and CRC, we undertook a large-scale meta-analysis of 63 studies, which included a total of 187,181 subjects . The results of our meta-analysis revealed that the C allele of rs4464148 , the T allele of rs4939827  and the T allele of rs12953717  were all associated with the increased CRC risk. Subgroup analysis according to ethnicity showed rs4464148 and rs12953717 were associated with the risk of CRC in both Caucasians and Asians, whereas rs4939827 was a risk polymorphism for CRC specifically in Caucasians. In summary, this large-scale meta-analysis indicated that SMAD7 polymorphisms  correlate with CRC.Mothers against decapentaplegic homolog 7 ( SMAD7 inhibits TGF-β signaling by preventing the formation of the SMAD2/SMAD4 complex [Cancer is caused by the dysfunction of intricate signaling pathways, leading to abnormal growth, metastasis, and many other events . The tra complex . It also complex .SMAD7 has also been reported to affect tumorigenesis via several other mechanisms. First, in FET-1 colon cancer cells, SMAD7 induces the expression of IκB, thereby repressing NF-κB activity [SMAD7 up-regulates MYC expression and WNT signaling via interactions with β-catenin in breast cancer [SMAD7 inhibits ERK1/2, JNK1/2, and p38 MAPKs under some circumstances related with tumorigenesis, such as erythroid differentiation [t cancer  and hepat cancer . In addintiation  and chonntiation .SMAD7 . Furthermore, they found these SMAD7 polymorphisms were associated with CRC adenomas and carcinomas [SMAD7 polymorphisms have been associated with the risk of developing multiple cancers, including CRC [SMAD7 polymorphisms and CRC risk.In 2007, Broderick and co-workers  conducteding CRC , and livding CRC . Howeverding CRC , breast ding CRC . These iFigure SMAD7 polymorphisms  in selected studies are shown in Table SMAD7 polymorphisms and CRC risk. Among these studies, one was conducted on rs12953717, with a relatively small sample size (308 subjects) [SMAD7 polymorphisms. Each subpopulation in the literature was treated as a separate study in our meta-analysis. Populations were divided into ethnic categories. The Newcastle-Ottawa Scale (NOS) was used for quality assessment [The characteristics of ubjects) , which ssessment  and all SMAD7 rs4464148 polymorphism and CRC risk, assuming different inheritance models. When all eligible studies were pooled into the meta-analysis, significant associations were found for the recessive genetic model . Subgroup analysis according to ethnicity showed that rs4464148 was significantly associated with CRC risk in both Caucasian and Asian populations  and the dominant genetic models: TT+ TC vs. CC , whereas it had no association with CRC risk among Asians : TT vs. TC+CC  and the dominant genetic models: TT+TC vs. CC , rs4939827 allele T vs. allele C , or rs12953717 allele T vs. allele C . The shape of Begg's funnel plot did not reveal any obvious asymmetry  has not yet been interpreted, a number of published epidemiological studies have reported that these polymorphisms are correlated with the risk of developing multiple cancers [TGF-β signaling is essential for maintaining homeostasis, cell differentiation, and tumor suppression , 24, 25.gression . Though  cancers , 29. How cancers .SMAD7 polymorphisms are related to risk of developing CRC. Our group has already used meta-analysis to systematically investigate the association between cancer risk and several SNPs involved in TGF-β signaling [SMAD7 polymorphisms  in the combined population were all significantly associated with CRC risk. Subgroup analysis according to ethnicity showed that rs4464148 and rs12953717 were significantly associated with the risk of CRC among both Caucasian and Asian population, whereas rs4939827 seems to be a risk polymorphism for CRC only within a Caucasian population. There could be several possibilities to explain such a differential association. First, the difference in association may result from differences in socioeconomic environment, regional dietary habits, and race. Second, the number of rs4939827 in Asian studies is still not as large as desired. In addition, the results from nine studies incorporated in this meta-analysis conflict with each other [SMAD7 rs4939827 and CRC in this ethnic group.These conflicting studies based their conclusions on small numbers of samples and different detection methods. Therefore ameta-analysis from large-scale samples of all available studies is required to have a more accurate assessment as to whether the ignaling –32. In tch other , 33–38. SMAD7 polymorphisms and CRC risk under an allelic model, while we also analyzed under dominant and recessive models. Therefore, our updated meta-analysis at a much larger scale clearly provides a more credible and reliable assessment for the association between SMAD7 polymorphisms and the risk to develop CRC.One recent study  also assSMAD7 polymorphisms with CRC risk, because there were not sufficient studies conducted on other cancers.Nonetheless, we also wish to acknowledge the limitations in our study. First, we stratified the studies by ethnic subtypes as Caucasian and Asian. However, we could not assess the association in the African population due to the insufficient number of African studies. Second, further subtle adjusted analysis could be carried out if more detailed individual information was available. Third, we only assessed the association of SMAD7 polymorphisms and cancer. However, controversies remain as whether those polymorphisms indeed associate with increased cancer risks. Our large-scale meta-analysis demonstrated that the C allele of rs4464148, the T allele of rs4939827, and the T allele of rs12953717 were all significantly associated with the increased CRC risk, which may provide a basis for genetic testing in the development of CRC. Consistent with our findings, Noci et al. [SMAD7 rs4939827 is also associated with cancer survival rate after therapy. Therefore, the identification of SMAD7 polymorphisms may also benefit developing targeted and personalized therapy against CRC. However, more comparative studies are needed to evaluate interactions of SMAD7 polymorphisms and cancer risk in other specific cancer subtypes and ethnic subtypesTo date, a large number of studies have focused on the relationship between i et al.  recentlySMAD7”, “Mothers against decapentaplegic homolog 7”, “rs4464148”, “rs4939827”, “rs12953717”, “polymorphism” or “variation”, “susceptibility”, and “tumor” or “cancer” or “carcinoma” or “neoplasia” or “colorectal caner” or “CRC” in PubMed, the Web of Science, EBSCO, and Embase databases. There were no limitations on the language and year for the literature search. The last search was updated on June 30, 2016. References of the retrieved publications were also screened.We searched for relevant case-control studies using the following words and terms: “SMAD7 polymorphisms  and CRC risk; (c) had available genotype frequencies both in cases and controls; (d) the genotype distribution in control groups was in Hardy-Weinberg equilibrium (HWE). (e) In cases of multiple studies with overlapping, redundant data published, only the most recent or complete study was included.Two authors independently screened titles and abstracts to identify relevant studies. Full-text articles of these studies were then carefully read to select eligible studies. Studies had to meet the following inclusion criteria: (a) were a case-control study, nested case-control or a cohort study; (b) evaluated the association between Two authors independently conducted the quality assessment. The Newcastle–Ottawa Scale (NOS) was used to evaluate the study quality, which scored studies by the selection of the groups, the comparability of cases and controls, and the ascertainment of the exposure. We considered a study awarded 0-3, 4-6, or 7-9 as a low-, moderate-, or high-quality study, respectively .P values according to the HWE in controls were extracted from the included studies.Two authors independently selected the relevant articles and extracted the following data: first author's name, publication date, ethnicity, cancer type, genotyping methods, number of cases and controls, and number of genotypes in case-control groups. In addition, SMAD7 polymorphisms and CRC by odds ratio (OR) corresponding to 95% confidence interval (CI) for different genetic models. Then we stratified the studies by ethnic subtypes and examined the association between SMAD7 polymorphisms and the CRC risk . Funnel plots were used to access publication bias by the method of Begg's test and Egger's test. A T test was performed to determine the significance of the asymmetry. An asymmetric plot suggested possible publication bias (EP ≥ 0.05 suggests no bias).Our meta-analysis was performed using Stata software . We first calculated the strength of the association between tic test  was perfed “high”. The fix


SOX17 , CDO1 , ZFP42  and TAC1 . Conclusions: The present meta-analysis demonstrates that methylated SOX17, CDO1, ZFP42, TAC1, FAM19A4, FHIT, MGMT, p16, and RASSF1A are potential superior biomarkers for the screening and auxiliary detection of lung cancer.Background: DNA methylation in sputum has been an attractive candidate biomarker for the non-invasive screening and detection of lung cancer. Materials and Methods: Databases including PubMed, Ovid, Cochrane library, Web of Science databases, Chinese Biological Medicine (CBM), Chinese National Knowledge Infrastructure (CNKI), Wanfang, Vip Databases and Google Scholar were searched to collect the diagnostic trials on aberrant DNA methylation in the screening and detection of lung cancer published until 1 December 2016. Indirect comparison meta-analysis was used to evaluate the diagnostic value of the included candidate genes. Results: The systematic literature search yielded a total of 33 studies including a total of 4801 subjects (2238 patients with lung cancer and 2563 controls) and covering 32 genes. We identified that methylated genes in sputum samples for the early screening and auxiliary detection of lung cancer yielded an overall sensitivity of 0.46 (0.41–0.50) and specificity of 0.83 (0.80–0.86). Combined indirect comparisons identified the superior gene of  Lung cancer is the leading cause of malignant tumor death, with the morbidity and mortality of lung cancer gradually increasing over the past decades . Only 13p16 [RASSF1A [APC [MGMT [DAPK [RARβ [With the emergence and development of molecular epidemiology, which opens the “black box” of the disease process, molecular biomarkers have much potential to improve the understanding of the occurrence, development, and prognosis of disease for the early detection of these lesions at the pre-invasive stage, even predicting the progress of the disease ,5. Nowadp16 , RASSF1A[RASSF1A , APC [10PC [MGMT , DAPK [1MT [DAPK  and RARβPK [RARβ . Many stPK [RARβ . The devPK [RARβ ,16 have However, the DNA methylation of specific genes in sputum samples do not yet provide a convincing and superior gene panel with high sensitivity and specificity for clinical utility in lung cancer. In addition, there was still no comparative evaluation on the diagnostic accuracy of methylation in sputum samples for the screening and detection of lung cancer. Recently, network meta-analysis has been developed to assess the comparative effectiveness of several interventions and to synthesize evidence across a network of studies ,18. ThisWe followed the reporting guidelines of the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) statement when conducting this review. The PRISMA statement has guidelines that include a four-phase flow diagram to systematically guide the inclusion and exclusion of research papers . In addiWe conducted a comprehensive literature search in PubMed, Ovid, Cochrane library, Web of Science databases, Chinese Biological Medicine (CBM), Chinese National Knowledge Infrastructure (CNKI), Wanfang, Vip Databases, and Google Scholar. The main search terms included: lung cancer or lung carcinoma or non-small cell lung cancer or “NSCLC”, “sputum or flema”, “diagnostic”, “sensitivity and specificity” and “methylation or hypermethylation or hypomethylation or demethylation”. All articles published until December 2016 were considered. In addition, the reference lists of all identified studies were manually searched to identify any additional studies.The articles, which could not be excluded based on the title and abstract, were retrieved for full-text review. Studies were included in this meta-analysis if they met the following criteria: (1) the diagnostic potential of sputum DNA methylation for lung cancer; (2) study design being case-control; (3) the patients being diagnosed with lung cancer by pathology; (4) provided data on the numbers of true positives (TP), false positives (FP), true negatives (TN), and false negatives (FN); (5) the methods of detecting methylation based on methylation-specific polymerase chain reaction (MSP) or quantitative methylation-specific PCR (qMSP).Studies were excluded from the meta-analysis for the following reasons: (1) abstracts, letters, reviews, expert opinions, case reports, or nonclinical studies; (2) studies had duplicate or overlapping data; (3) study was based on tissue, blood, or animals.Data extraction was conducted in duplicate by two investigators (Di Liu and Hongli Peng) based on title, abstract, author, year of publication, country of origin, sample size, assay methods, and diagnostic performance (sensitivity (SEN), specificity (SPE), TP, FP, FN, TN), target gene(s), and the score of the quality assessment of studies of diagnostic accuracy (QUADAS)  and the STARD and QUADAS guidelines were utilized to assess the methodological quality of each study. There are 25 items in the STARD initiative checklist, and a score of 1 was given when the item was yielded . FourteeI2 statistic, which measures the extent of inconsistency between studies, was also assessed [We used standard methods recommended for the direct meta-analysis which estimated the diagnostic test evaluation of DNA methylation compared with the gold standard . The numassessed . Spearmaassessed . Analysep-value of less than 0.05 was considered significant.For indirect comparisons, the comparative diagnostic accuracy of all biomarkers was estimated according to common comparator (the gold standard). We did not assume consistency  of two biomarkers without direct analyses. We took the step-wise approach, which was suitable for the simple star-network meta-analysis to obtain an indirect analysis. The Deeks’ test and Egger’s test were utilized to estimate the funnel plot asymmetry and the publication bias . The indP16, RASSF1A, APC, MGMT, PAX5, CGB, GATA, DAPK, RARβ, MAGE, HOXA, 3OST2, PRDM14, FAM19A4, PHACTR3, PCDH20, CXCL, Dal-1, Dab2, Dcr2, SULF2, Kifla, Jph3, SOX17, CDO1, ZFP42, TAC1, CDH1, H-cadherin, FHIT, RASSF2, TCF2l).The flowchart of included studies was presented in ZFP42: 0.87 (0.78–0.93), TAC1: 0.86 (0.77–0.92), SOX17: 0.84 (0.75–0.91), FAM19A4: 0.80 (0.74–0.85), and HOXA: 0.79 (0.63–0.89); whereas the higher of PLR were SOX17: 6.76 (2.34–19.54), FHIT: 5.93 (2.29–15.36), RASSF1A: 5.61 (3.73–8.43), MGMT: 4.78 (1.47–15.55), and p16: 4.71 (2.53–8.78).As shown in The graph of the SROC (summary receiver operating characteristic) cure is shown in SOX17, TAC1, ZFP42, CDO1 differed significantly from the other 28 methylated genes and were higher than 1, indicating that SOX17, TAC1, ZFP42, and CDO1 have a higher diagnostic accuracy. More information is shown in the SOX17 , CDO1 , ZFP42 , TAC1  (y: 0.75) .p < 0.001), indicating the heterogeneity of threshold effect. We also investigated the non-threshold effects; the results indicated the existence of significant heterogeneity in the overall sensitivity , specificity , PLR , NLR , and DOR . Therefore, a bivariate binomial mixed model was applied to summarize the pooled estimates in this study. To determine the sources of heterogeneity, we performed meta-regression to test the effect of ethnicity (Asian/others), sample size (n = 0–100/101–200/201–), and the quality of study (low/medium/high). Multivariable regression showed that ethnicity  and the sample size  had statistically significant differences, while the quality of study  showed no significant difference. Then, we conducted subgroup analysis based on ethnicity and the sample size, as shown in In the meta-analysis, computation of the Spearman’s rank correlation coefficient between the logit of sensitivity and that of 1-specificity of sputum DNA testing was 0.465 (p = 0.048) and the sample size  were of statistically significant bias for p16, while not significant bias for other candidate genes (p > 0.050).In addition, we performed a meta-regression to test the effect of ethnicity, sample size, and the quality of study with different genes. The results showed that ethnicity , CDO1 , ZFP42 , and TAC1 . A single DNA methylation biomarker cannot be expected to detect all cases of lung cancer. Some studies demonstrated that combined multiple methylated genes could improve the diagnostic value of cancers [FAM19A4 and PLR value of methylated RASSF1A, FHIT, MGMT, and p16 are relatively high, suggesting that they are comprehensive parameters for the screening test [RASSF1A and p16 genes are reported to be promising driving molecules in many cancers under the concept of precision medicine [FAM19A4, FHIT, and MGMT were reported to play important roles in the occurrence and deterioration of lung cancer [SOX17, CDO1, ZFP42, TAC1, FAM19A4, FHIT, MGMT, p16, and RASSF1A are useful in the screening and auxiliary detection of lung cancer.In the direct meta-analysis, we identified that methylated genes in sputum samples for the early screening and auxiliary detection of lung cancer yielded an overall sensitivity of 0.46 at the same specificity of 0.83. Furthermore, the PLR (positive likelihood ratio) was 2.72, NLR (negative likelihood ratio) was 0.64, and DOR (diagnostic odds ratio) value was 4.28, and the AUC (area under the curve) was 0.69, indicating a mid-level accuracy. Therefore, we should pick the superior genes for clinical utility as diagnostic biomarkers for lung cancer. Combined indirect comparisons identified the superior genes as  cancers ,60. We iing test . In addimedicine ,66,67. Ig cancer ,69,70,71To our knowledge, this study is the first systemic review and indirect meta-analysis to assess the comparative diagnostic effectiveness of the methylation profile of multiple candidate gens in sputum samples for the early screening and detection of lung cancer. According to the method of network meta-analysis ,27, we up = 0.001) and the sample size (p = 0.036) might be a source of heterogeneity for this meta-analysis. The results of subgroup analyses showed that large-sample studies had higher sensitivity than the small- and moderate-sample studies, while studies in Asia had lower sensitivity than other regions. However, we observed a large degree of heterogeneity among studies investigating methylation profile in sputum samples used for lung cancer. Threshold effect is one of the primary causes of heterogeneity among diagnostic accuracy studies . In the In addition, publication bias could exist in the meta-analysis. This meta-analysis was only based on published studies, therefore inducing the possibility of publication bias. The Deeks’ test and Egger’s test not only detect publication bias, but also indicate the heterogeneity due to the effect of ethnicity, sample size, the quality of study, etc. ,73. TherThe present network meta-analysis included 33 articles and 32 candidate genes, with the majority of genes only included in one article. These make it difficult to directly compare the diagnostic efficacy among multiple genes; thus, only indirect comparisons were evaluated in this study. However, the absence of direct comparisons may lead to bias. Pairwise meta-analysis and network meta-analysis were carried out sequentially for direct and indirect comparisons of migraine headache days among three interventions compared with those treated by three placebos, and the results showed that there was no significant inconsistency between the direct and indirect evidence for the majority of comparisons . AnotherBased on the focus of diagnostic accuracy studies, we identified other common limitations and insufficiency. Firstly, all the publications included in this analysis were reported on case-control studies, indicating that the selection bias could possibly lead to over-estimations of diagnostic accuracy compared with the cross-sectional study and cohort study ,77. In aSOX17, CDO1, ZFP42, TAC1, FAM19A4, FHIT, MGMT, p16, and RASSF1A are useful biomarkers in the screening and auxiliary detection of lung cancer. Our findings provide new avenues for assessing the comparative diagnostic effectiveness of several methylations in lung cancer based on the method of network meta-analysis. Further high-quality and large-scale studies are needed to confirm our analysis.In conclusion, despite these limitations, our meta-analysis advocates that methylated


Data on the association between aspirin and statin use and Pancreatic Ductal AdenoCarcinoma (PDAC) risk are conflicting. These drugs are often co-prescribed, but no studies evaluated the potential combined or confounding effect of the two at the same time. We aimed to investigate the association between aspirin and statin exclusive and combined use and PDAC occurrence. Data on environmental factors, family and medical history were screened in a case-control study. PDAC cases were matched to controls for age and gender. Power calculation performed ahead. Odds ratios (OR) and 95% confidence intervals(CI) were obtained from multivariable logistic regression analysis. In 408 PDAC patients and 816 matched controls, overall statin , but not aspirin use was associated to reduced PDAC risk. Compared to non-users, exclusive statin  and exclusive aspirin users  had reduced PDAC risk. Concomitant statin and aspirin use did not further reduce the risk compared with statin use alone and no interaction was evident. Statin protective association was dose-dependent, and consistent in most subgroups, being stronger in smokers, elderly, obese and non-diabetic patients. The present study suggests that statin use is associated to reduced PDAC risk, supporting a chemopreventive action of statins on PDAC. PDAC 5-year survival rate is only 6%1 as only 20% of patients are eligible for surgery3, and chemo/radiotherapy marginally improve survival in advanced disease4. Screening is advised only for high risk individuals in experimental settings5.Pancreatic ductal adenocarcinoma (PDAC) is the 46. Evidence regarding potential chemopreventive drugs for PDAC is instead limited.Prevention might, therefore, play a key role in terms of modifiable lifestyle risk factors such as smoking and/or alcohol drinking, overweight, obesity and diet7 particularly colorectal cancer (CRC)9.Aspirin is a nonsteroidal anti-inflammatory drug (NSAID), irreversible inhibitor of the cyclooxygenase 1/2, widely prescribed for its anti-inflammatory, antipyretic, analgesic and anti-platelet activities. Its chemopreventive role is well recognized for many cancer types12. However, its role in clinical studies on PDAC is less clear, and the most recent meta-analysis on cohort and case-control studies, showed an overall protective effect of aspirin with an odds ratio (OR) of 0.77, although with high heterogeneity13.Preclinical studies demonstrated that aspirin has a role in inhibiting different signaling pathways involved in the carcinogenesis of several cancers including PDAC, such as mTOR, NFkB and Wnt, resulting in enhanced DNA mismatch repair mechanisms, apoptosis, angiogenesis and tumour progression inhibition14. The mevalonate pathway involved in their action has a role on multiple signaling cascades such as Ras, MEK, mTOR, BCL-2 and Rho kinases17, that play role in carcinogenesis and tumour progression. Statin antineoplastic effect is also related with an immunomodulatory and anti-inflammatory activity and with angiogenesis inhibition19. Their clinical role as chemopreventive agents is still controversial, although a few meta-analyses showed a risk reduction associated with their use, mostly for gastrointestinal tract tumours22. The latest meta-analysis on the association between statin use and PDAC risk showed no pooled effect23. However, since then, other large case-control studies were published25, recording a reduced PDAC risk associated with statin use.Statins are 3-hydroxy-3-methyl-glutaryl-coenzyme A (HMG-CoA) reductase inhibitors commonly prescribed for the treatment of hypercholesterolemia and for cardiovascular primary and secondary preventionvice versa. An investigation on the use of both these drugs, alone and in combination, has been conducted by Hoffmeister et al. in a case-control study on CRC. The authors reported a stronger risk reduction for statin exclusive use than for that of aspirin, while the combination of the two drugs did not further increase the effect of statins, unless for prolonged uses26.As aspirin and statins are often co-prescribed, mostly for cardiovascular disease prevention or treatment, one might hypothesize that in studies investigating aspirin, at least a part of the observed effect might be due to the use of statins and As studies on the possible chemopreventive activity of both aspirin and statins on PDAC gathered heterogeneous results, the analysis of their potential combined or confounding effect in this tumour type seems particularly important. However, there are no studies examining the association of both these two drugs at the same time with PDAC risk.Therefore, the primary aim of the study was to examine the association between overall and exclusive aspirin and statin use and their combined use with the risk of PDAC. In addition, we conducted subanalyses to explore whether the association was stronger in specific subgroups.A single-center case-control study was conducted from January 2006 to February 2016. Patients enrolled in either group provided written informed consent for interviews. The study received local IRB approval at Sant’Andrea Hospital. Methods were performed in accordance with the relevant guidelines and regulations.Incident cases were prospectively recruited at the gastroenterology department and had PDAC histological diagnosis. Controls consisted of hospital non-patient visitors not genetically related to cases, as well as outpatients and inpatients from the gastroenterology department. Both cases and controls demonstrated the will and ability to participate providing personal data, clinical and cancer history. Exclusion criteria for controls in order to reduce the risk of recruitment bias were: (a) personal history of inflammatory bowel disease, chronic kidney disease or liver cirrhosis, (b) referral to our center for family history (FH) of gastrointestinal cancer, (c) referral to our center or hospitalization for NSAID-induced ulcer disease, (d) personal history of neoplasia in the last 5 years.We used a frequency matching oversampling controls with a 2:1 ratio. For each case enrolled, the first two eligible controls of the same sex and age (±2 years) were enrolled and interviewed within 30 days.Data were recorded on a standardized form by a trained physician through direct patient interview; no proxies were interviewed.The following clinical, epidemiological, therapeutic and morphological parameters were collected: sex, age, race, tobacco and alcohol intake, body mass index (BMI), FH of cancer, previous pancreatic diseases, history of diabetes, aspirin and statin use, length, type and dosage of their use. During the interview, a list of brand and generic medication names for aspirin and statins was provided to help facilitate recall. All cases were interviewed within 1 month from diagnosis and data pertaining the disease were also recorded.For ever smokers, a consumption of at least 100 cigarettes or >6 months of smoking were needed to be considered a smoker. The total amount of smoking was evaluated as pack-years, defined as the product of packs smoked per day and the total years of smoking. A cut-off of 20 pack-years was set to define heavy smokers. Current smokers were considered as smokers who were currently smoking or who had quit less than 1 year in the past. Ex-smokers were classified as smokers who had quit at least 1 year before the diagnosis of the disease, or its first presentation symptom for cases, or the time of interview for controls.For ever drinkers, a consumption of at least 12.5 g (1 unit) of alcohol/month was needed to be considered a drinker. One glass of wine, 1 pint or can of beer, one shot of hard liquor were each considered approximatively equal to 12.5 g of alcohol. A cut-off of 21 units/week  was set to define heavy drinkers.28.This cut-off was used because in a meta-analysis of 156 studies, drinking was reported as a risk factor for many cancers at a dose of both 25 and 50 g/day, and population studies in Italy suggest that 94% of subjects report either being teetotallers or consuming ≤4 alcoholic drinks daily2 (kg/m2) with obesity considered as BMI >30 kg/m2. Diabetes was recorded as a potential risk factor when diagnosed >1 year before the diagnosis of cancer or its first symptoms for cases or before the interview for controls.BMI was calculated as usual adult weight/heightAspirin or statin use was defined as the ever use of the medication for at least three consecutive months.29. Statin users were categorized in “low-dosage users” (<20 mg) and “moderate/high dosage users” (≥20 mg) based on median value.Aspirin users were categorized in “high dosage users” (≥300 mg) and “low dosage users” (≤160 mg)Aspirin or statin users were categorized into different length of duration  based on median value.To avoid possible bias due to cancer symptoms  or subsequent cancer therapies, subjects were asked about risk factors that were present 12 months before diagnosis or presentation symptoms.A power calculation was performed ahead: considering an exposure of ~20% for aspirin or statins as previously recorded in 200 controls, to have a 80% power of identifying an odds ratio (OR) ≤ 0.62 as single effect of aspirin or statins, 395 cases and 790 controls were needed. This number would also allow to detect an OR ≤ 0.50 for the combined used of aspirin and statins, based on a 10% combined exposure among controls.A descriptive analysis was conducted to show the characteristics of PDAC patients at time of diagnosis. Case-control comparisons were made using Chi-square and Fisher’s exact tests where appropriate, for categorical variables and Student’s t-test for continuous variables.Logistic regression was used to calculate ORs and their 95% confidence intervals (CI). Multivariable logistic regression models were adjusted for potential confounders: age at diagnosis (for cases) or interview (for controls) (5-year age groups), gender, BMI, smoking history, drinking habits, diabetes history, chronic pancreatitis history, FH of PDAC.P values are 2-sided. P values < 0.05 were considered statistically significant.All statistical analysis were performed using MedCalc version 13 . All reported The STrengthening the Reporting of OBservational studies in Epidemiology (STROBE) checklist for case-control studies was checked for items that should be included in the report.Of 421 PDAC cases seen at the Gastroenterology Department during the study period, 9  were not histologically confirmed. Of the remaining 412, 2 (0.5%) refused to participate and 2 (0.5%) were too ill to take part; of the 893 controls recruited in the same timeframe, 51 (5.7%) were excluded for matching exclusion criteria, 8 (0.9%) refused to be interviewed for privacy concerns, 18 (2.0%) were too ill to take part. This led to a participation rate of 99% for cases and 91% for controls; the analysis was therefore conducted on a final population of 408 cases and 816 matched controls. Respectively 48.7%, 13.4% and 37.9% of the controls, consisted of visitors, inpatients and outpatients. Most of the inpatients were hospitalized for diverticulitis or gastrointestinal infections; outpatients were visiting for either gastro-esophageal reflux disease, irritable bowel syndrome, chronic constipation or dyspepsia.The median age of cases and matched controls was 68 years (range 35 to 99); 51.2% were men.Almost all cases and controls were Caucasians.st degree FH of PDAC, previous history of diabetes, previous chronic pancreatitis, and were more frequently smokers, heavy smokers and heavy drinkers  and 191 controls 23.4%) reported ever use of aspirin, which was associated with a statistically borderline significant reduced PDAC risk  and 203 controls (24.9%) reported ever use of statins, which was associated with a statistically significant reduced PDAC risk  but not aspirin  use was associated to a reduced PDAC risk Table . OR and In order to avoid possible bias due to controls selection, a sensitive analysis for control type  was performed. At multivariable analysis, statin use was associated to a reduced risk of PDAC both with sensitive analysis restricted to visitors  or to hospital controls , while the use of aspirin was not associated to PDAC risk for either visitors  nor hospital controls .vice versa  was associated to a reduced risk of PDAC occurrence, while the association for aspirin use was of borderline significance . Also at multivariable analysis the combined use of the two drugs did no further reduce the risk compared to the use of statins alone . We found no evidence of interaction between statin and aspirin (P = 0.17), when adding an interaction term to the main effects of aspirin and statin in a multivariable model adjusted for other potential confounding variables  as expected, an increased PDAC risk for patients with multiple 1st degree FH of neoplasia, 1st degree FH of PDAC, increased BMI, previous history of diabetes, chronic pancreatitis, smoking habit was seen, suggesting the absence of biases and the genuineness of our population; 4) statin results are consistent with most Bradford Hill criteria for causation44 as the association is strong, a biological gradient is evident as higher dosages lower the risk and the association is plausible in terms of mechanisms and coherent with in vitro studies. A clear temporal relation was instead not evident. Of course, as this is a case-control study, causation of an effect cannot be observed and we can only report an association with risk, useful to generate hypotheses that need validation.The present study has some strengths: 1) it is the first specifically aimed at evaluating the association between overall, exclusive and combined use of both aspirin and statins and PDAC risk, as they both seem to have chemopreventive properties and are very often prescribed together; 2) it is one of the few studies on this topic with an 45. To explore the possibility of a potential selection bias among hospital controls, who might have been less likely to use aspirin being selected in a Gastroenterology setting, we performed a sensitivity analysis comparing the effect of aspirin and statins in the two different control groups  separately with findings suggesting a consistent effect of statins but not of aspirin.There are, however, some limitations. First, as for any case-control study, the risk of recall bias has to be taken into account, although when interviewed, patients were provided with a list of brand and dosages in order to reduce this risk. Furthermore, both the rate of non-participants and the rate of patients not recalling data about their drug use are similar among cases and controls. In order to reduce the risk of confounding factors, the questionnaire was carefully conducted by a trained physician with expertise on pancreatic disorders, asking information regarding risk factors exposure as present 1 year before the interview. As in any case-control study the choice of the control population is a possible matter of concern. We opted for a control group that we believed to represent the same population as the case group, derived from the same catchment area, and being composed both of patients seen for similar symptoms by the same doctors in the same clinics but with a final diagnosis unrelated with the disease of interest, and by hospital non-patient visitors; this, therefore, was unlikely to cause a specific bias47 and to populations used for other case-control studies on PDAC48. Also, as the study was not powered for the exclusive use analyses or subgroup analysis, these results have to be taken carefully into account as they might not be conclusive. For example, in a post-hoc calculation limited to the 653 smokers (253 cases and 400 controls), the statistical power of the study dropped to 67% for detecting an OR = 0.62 (as per protocol) considering the actual aspirin exposure (26%) measured among control smokers.Furthermore, our control group seems to be representative of the general Italian population in terms of environmental risk factors and exposure to the drugs of interest49.Furthermore, one cannot exclude that the observed reduced PDAC risk associated with statin use and to a lesser extent of aspirin, is a surrogate for other uninvestigated factors such as a healthier lifestyle. At any rate, the current analyses were corrected for most known risk factors associated with PDAC risk 53 might be of interest.The biological mechanisms through which statins might prevent PDAC are not completely clarified. Interestingly, the importance of statins has been recently proved in retrospective studies also in terms of prolonging survival in operated PDAC patientsIn conclusion, the present study suggests that statin use, rather than that of aspirin, in particular at higher dosages is associated to a reduced PDAC risk. These findings support a chemopreventive action of statins on PDAC and no apparent synergistic activity of the two medications.Supplementary Tables


However, the conversion patterns of FDG-PET metabolism across studies are not conclusive. We conducted a voxel-wise meta-analysis using Seed-based d Mapping that included 10 baseline voxel-wise FDG-PET comparisons between 93 aMCI converters and 129 aMCI non-converters from nine longitudinal studies. The most robust and reliable metabolic alterations that predicted conversion from aMCI to AD were localized in the left posterior cingulate cortex (PCC)/precuneus. Furthermore, meta-regression analyses indicated that baseline mean age and severity of cognitive impairment, and follow-up duration were significant moderators for metabolic alterations in aMCI converters. Our study revealed hypometabolism in the left PCC/precuneus as an early feature in the development of AD. This finding has important implications in understanding the neural substrates for AD conversion and could serve as a potential imaging biomarker for early detection of AD as well as for tracking disease progression at the predementia stage.Brain The online version of this article (10.1186/s40035-018-0114-z) contains supplementary material, which is available to authorized users. Alzheimer’s dementia (AD), an age-related neurodegenerative disorder, is the most common type of dementia worldwide, placing heavy burdens on global economic and social development –3. Mild 18F-fluorodeoxyglucose (FDG) PET is now a well-established tool in detecting and defining the distribution of neural injury or synaptic dysfunction in AD and MCI/aMCI [In contrast to brain amyloid-beta positron emission tomography (PET) that has no place in the routine work-up of suspected AD , brain 1MCI/aMCI –9. CerebMCI/aMCI , 10. A sMCI/aMCI . Many loMCI/aMCI , 13. HowMCI/aMCI –18, PCC MCI/aMCI –20, precMCI/aMCI , medial MCI/aMCI , medial MCI/aMCI , 20, 23.MCI/aMCI . HoweverMCI/aMCI . In addiMCI/aMCI , 21, witMCI/aMCI , 23. FurMCI/aMCI –27. TherMCI/aMCI .d Mapping (SDM), a popular meta-analytical technique [The aim of this work was to carry out a modified meta-analysis in order to identify brain regions of FDG-PET metabolic alterations that could be used to predict the conversion from aMCI to AD. In order to do this, we pooled voxel-based FDG-PET studies to determine the most consistent and reliable brain regions that showed alterations in glucose metabolism. The meta-analysis was conducted using Seed-based echnique , 28–31 technique , 32–36.Electronic searches were conducted in the PubMed, Web of Science, and Embase databases using a combination of keywords (“Mild cognitive impairment” OR “Alzheimer’s disease”) AND ((“Positron Emission Tomography” OR “PET”) AND (“Fluorodeoxyglucose” OR “FDG”)) OR “FDG-PET”) on February 17, 2017. No publication language or date restrictions were used for the electronic searches. The reference lists of relevant original studies, systematic reviews, and meta-analyses were manually screened for additional qualified publications as well. Criteria for considering studies for this meta-analysis were as follows: (1) studies were published as original research articles in peer-reviewed journals; (2) studies were longitudinal that followed patients diagnosed with aMCI (single domain or multiple domains) –41 to dehttps://www.sdmproject.com/). The SDM approach has been described in detail in other publications [http://www.sdmproject.com/software/tutorial.pdf). Briefly, an effect-size signed map and an effect-size variance map from the peak coordinates and their effect sizes of regional glucose metabolic differences between aMCI converters and aMCI non-converters was first separately recreated for each individual FDG-PET study using an anisotropic via a 20 mm full-width at half maximum (FWHM) un-normalized Gaussian kernel. It should be noted that this kernel is designed to assign indicators of proximity to report coordinates but not to smooth any image, which is different in nature from the smoothing kernel. Moreover, this kernel has been found to have excellent control over false positives. Following this, a mean map was created by calculating the random-effects mean of the study maps in a voxel-wise manner, which was weighted by sample size and intra-study variability, and additional between-study heterogeneity. Statistical significance was determined using a default threshold , which optimized the balance of false positives and negatives [Regional glucose metabolism differences between aMCI converters and aMCI non-converters were meta-analyzed in a voxel-wise manner using the SDM software package  [In order to assess the replicability of the results, a whole-brain voxel-based jackknife sensitivity analysis was performed by iteratively repeating the same statistical analysis, removing one study at a time , 29. Jac voxels) , 28.A random-effects model with Q statistics  was utilized to examine the between-study heterogeneity of individual clusters. A permutation approach  was used to determine statistical significance .p-value less than 0.05 on Egger’s test.Possible publication biases in the main meta-analytic peaks were examined by analyzing the funnel plots and Egger’s tests . SignifiMeta-regression analyses were conducted to explore the confounding effects of baseline mean age and severity of cognitive impairment (mean MMSE score), and follow-up duration that could potentially influence the meta-analytic results. Statistical significance was thresholded at a more conservative p-value of 0.0005 and a cluster extent of 10 voxels in accordance with previous meta-analyses , 30.Our initial search yielded a total of 3886 titles reviewed for inclusion. After a careful screen of the literatures according to the inclusion and exclusion criteria, nine original studies that reported 10 comparisons –23, 45 wAs shown in Fig. The jackknife sensitivity analyses revealed that regional hypometabolism in the left/right PCC/precuneus  was the most reliable alteration as it was replicable in all 10 comparisons. Regions of hypometabolism in the left/right ACC  and left middle temporal gyrus  were replicable in all but one comparison combinations. Regions of hypometabolism in the left middle temporal gyrus (BAs 21 and 37) and left middle frontal gyrus, orbital part (BA 11) failed to emerge in two of the comparison combinations. These results are summarized in Table As demonstrated in Additional file No significant publication biases were detected in the regions obtained from the main voxel-wise meta-analysis as revealed by the relative symmetry of the funnel plots  exhibited more regional hypometabolism in the left middle temporal gyrus  and less regional hypometabolism in the left superior/middle frontal gyri, orbital part (BA 11). Lower mean MMSE scores of aMCI converters at baseline  were associated with more regional hypometabolism in the left middle frontal gyrus (BA 9) and left temporal pole (BAs 28 and 36). In addition, findings of the meta-regression analysis suggested that a longer mean follow-up duration in aMCI-converters  was associated with more regional hypometabolism in the left middle frontal gyrus (BAs 46 and 9). The results of these meta-regression analyses are summarized in Table To the best of our knowledge, this was the first voxel-wise meta-analysis that only pooled FDG-PET studies to determine the most robust brain regions with glucose metabolism alterations in prediction conversion from aMCI to AD. In addition to the main voxel-wise meta-analysis in this study, several complementary analyses, such as jackknife sensitivity, heterogeneity, and publication bias analyses were conducted to test the robustness of our results. Pooling these findings, we observed that regional hypometabolism in the left PCC/precuneus at baseline was the most reliable and robust difference between aMCI converters and aMCI non-converters. Moreover, meta-regression analyses revealed that regional hypometabolism in the left PCC/precuneus in aMCI converters relative to aMCI non-converters at baseline was not biased by potential confounding variables, such as mean age and severity of global cognitive impairment at baseline, or follow-up durations across studies, which lend further support for the robustness of this finding.The PCC and its adjacent precuneus, situated in the posteromedial part of the parietal cortex, make intensive connections with the medial temporal lobe (MTL) and other cerebral areas –48. The Interestingly, in terms of consistency of prediction, we observed an asymmetric pattern of regional hypometabolism. Although in the main voxel-wise meta-analysis, the right PCC/precuneus was identified, these regions were associated with significant between-study heterogeneity. Including more eligible studies in future investigations may increase our statistical power. The underlying neurobiology of the asymmetric involvement remains unclear; however, increasing evidence indicates greater susceptibility of the left hemisphere to degeneration than the right. Moreover, it has been suggested that brain structures located in the left hemisphere are affected earlier and more severely in AD –65. IndeThe findings of the present meta-analysis were not consistent with those of a prior meta-analysis that combined three different imaging modalities  . That meIn addition to the left PCC/precuneus identified in the main voxel-wise meta-analysis, several other brain regions exhibited hypometabolism, such as the right PCC/precuneus, bilateral ACC, and left middle temporal and frontal gyri. However, the finding of altered glucose metabolism in these regions was not very reliable, as determined by additional complementary analyses of jackknife sensitivity, heterogeneity, publication bias, and meta-regression. Several moderator variables were examined to understand the potential source of this heterogeneity. The meta-regression analyses indicated that mean age and severity of cognitive impairment at baseline as well as follow-up duration had significant effects on alterations in brain metabolism. Many other moderators, such as age at onset, disease duration, education level, aMCI subtype (amnestic single domain and multiple domains), genotype, and cerebrovascular risk factors might also affect brain metabolism. In addition, variations in imaging protocols and analytic procedures across studies may have influenced the observed heterogeneity. However, we were unable to address many of the described confounding factors as we were limited to performing analyses on data available from the included studies. Future studies with large, homogeneous samples and standardized protocols are warranted to control these moderator effects and to verify the results of our meta-analysis.Although our comprehensively modified meta-analysis enabled us to successfully identify which brain regions might serve as predictive indicators of conversion from aMCI to AD, the present study had several limitations. First, the SDM technique was based on reported peak coordinates and their effect sizes rather than raw imaging data. Although this is common practice in coordinate-based meta-analyses, the approach may produce less accurate results , 70–72. Using a modified meta-analytic approach, the present study demonstrates the most robust FDG-PET changes in the left PCC/precuneus in aMCI converters relative to aMCI non-converters at baseline. This finding has important implications in understanding the neural substrates for the prediction of conversion from aMCI to AD. Hypometabolism in the left PCC/precuneus was determined as an early feature in the development of AD and this might serve as a candidate imaging biomarker for early detection and for tracing disease progression. Further work is warranted to determine the potential of these findings in clinical use and their value in the development of future therapeutic interventions in individuals at the predementia stage of AD.Additional file 1Figure S1. Regions of the heterogeneity map from the SDM heterogeneity analysis. Abbreviations: SDM, Seed-based d Mapping; L, left; R, right. The color bar indicates the maximum and the minimum SDM-Z values. (TIFF 684 kb)Additional file 2Figure S2. Funnel plots of the peak coordinates from the main meta-analysis for detecting publication bias. Abbreviations: MNI, Montreal Neurological Institute. (TIFF 184 kb)


P = 0.004). In the subgroup analysis of study design, both case-control studies and prospective cohort studies showed significant results , while retrospective cohort studies did not show positive result . In a further stratified analysis by type of CVD, a significant association was found between PCOS and coronary heart disease (CHD) . However, no significant association was observed between PCOS and myocardial infarction (MI) . In conclusion, this meta-analysis suggested that PCOS is significantly associated with increased CHD risk.Some studies reported a significant association between polycystic ovary syndrome (PCOS) and risk of cardiovascular disease (CVD). However, the results are controversial. A systematic search was conducted in the PubMed, Science Direct, EMBASE, and Cochrane Library databases. Five case-control studies and 5 cohort studies were selected, involving a total of 104392 subjects in this meta-analysis. PCOS was significantly associated with the increased risk of CVD (OR = 1.30; 95% CI 1.09 – 1.56;  Polycystic ovary syndrome (PCOS) is one of the most common cause of infertility in women population . Around Cardiovascular disease (CVD) is one of the most common cause of mortality worldwide. Previous studies suggested that there will be more than 23 million CVD patients by 2030 -6. The aA total of 121 potential studies were found by searching online databases, such as PubMed, Web of Science, Science Direct, EMBASE, and Cochrane Library. Five case-control studies and five cohort studies were selected in this study. A total of 104392 subjects were included in this meta-analysis. The detailed literature search strategy was showed in Figure P = 0.004). In the subgroup analysis of study design, both case-control studies and prospective cohort studies showed significant results , while retrospective cohort studies did not show positive result . In a further subgroup analysis by the type of CVD, PCOS patients had significantly increased risk of coronary heart disease (CHD) . However, PCOS patients did not have significantly increased risk of myocardial infarction (MI) . In addition, both large sample size studies and small sample size studies showed significant results . The results were showed in Table As shown in Figure 2 = 0%, P = 0.74). Furthermore, the result was not influenced . In addition, Begg's funnel plot .Sensitivity analysis suggested that the results of this study were statistically reliable Figure . As showThis is a meta-analysis which aimed to clarify the association between PCOS and the risk of CVD. Ten studies involving more than 100000 individuals were included in our meta-analysis. The results suggested that PCOS was a risk factor of the increased risk of CVD. In a stratified analysis by type of CVD, a significant association was found between PCOS and CHD. However, no significant association was observed between PCOS and MI. Two previous meta-analyses also found that women with PCOS appeared to be at increased risk of CHD , 18. ThuA recent meta-analysis determined the relationship between PCOS and CVD risk markers . That stThis study had some advantages. First, substantial number of individuals were pooled from different trials, which significantly increased statistical power of the meta-analysis. Second, the quality of studies included in current meta-analysis was relatively satisfactory and met our predefined inclusion criteria. Third, no publication bias was found, which suggested that the result was unbiased.Several limitations of this study must be acknowledged. Firstly, only online databases were searched; some studies may have been missed. Secondly, no studies with Asians and other races was included in this meta-analysis. Thirdly, heterogeneity was found in this study. Sensitivity analysis was used to find the source of heterogeneity. Fortunately, it was decreased when one outlier was excluded.This study suggested that PCOS patients had significantly increased CHD risk.PubMed, Science Direct, EMBASE, and Cochrane Library databases were searched by two independent authors (LQ Zhao and ZG Zhu). Last search was updated in Apr, 2016. The search terms were used as follows: (polycystic ovary syndrome or PCOS) and (cardiovascular diseases or CVD). No publication date or language restrictions were imposed. We contacted primary authors to clear doubts. The reference list of obtained articles is showed in the supplemental material.The selection criteria of the retrieved articles in our meta-analysis were as follows: a) case-control studies or cohort studies; b) studies evaluating the association between PCOS and risk of CVD; c) sufficient data available to calculate an odds ratio (OR) with 95% confidence interval (CI). The exclusion criteria of the meta-analysis were: a) case-only studies; b) studies with incomplete data; and c) meta-analyses, letters, reviews, and editorial articles. If more than one study was published by the same author using the same patient population or overlapping case series, the study with the largest size of samples was included.Two investigators extracted data from the included studies independently, and the respective studies were retrieved for further consideration if judged pertinent by one or two reviewers. Any discrepancies were identified and resolved by consensus. For each study, the following data were extracted: first author's name, year of publication, study design, race, age, duration of follow-up, sample size, outcomes, and covariant. A modification of the Newcastle-Ottawa Scale (NOS) was used as an assessment tool for selection, comparability, and outcome assessment. The studies should (1) brief description of appropriateness of studies assembled for assessing the hypothesis; (2) rationale for the selection and coding of data ; (3) documentation of how data were classified ; (4) assessment of confounding variables ; (5) assessment of heterogeneity.2 test statistic. For the Cochran's Q test statistic, a P value < 0.10 was accepted as statistically significant heterogeneity. Random-effects models were used to estimate summary ORs and 95% CIs. We also conducted subgroup analyses by study design, type of outcomes, and sample size. Galbraith plot was also performed to identify sources of heterogeneity. Sensitivity analyses were conducted to assess the strength of our findings by excluding one study at a time. Begg's funnel plot and Egger's regression test were used to evaluate publication bias. In Egger's test, when P value < 0.10, it was considered statistically significant publication bias. All analyses were conducted using Stata v.12  statistical software.The strength of the association between PCOS and CVD risk was estimated by calculating ORs with 95% CIs. Tests for heterogeneity were made among studies using the Cochran's Q and I