P = 0.105], CSS  and OS , by fixed or random effects models. The findings were consistent independently of age, sample size, publication year, follow-up duration, study type and geographical region. In summary, the present findings demonstrate that the presence of positive surgical margins is associated with poor survival outcomes in bladder cancer following RC, indicating that avoidance of positive surgical margins during surgery is helpful to improve the prognosis of patients with bladder cancer.Data regarding the association between surgical margin status and the outcome of bladder cancer treated by radical cystectomy (RC) are conflicting. Therefore, the present meta-analysis was performed to assess the associations between the outcomes of bladder cancer, in terms of recurrence-free survival (RFS), cancer-specific survival (CSS) and overall survival (OS), and the presence of positive surgical margins versus negative surgical margins following treatment with RC. Research articles published prior to April 2016 were identified from Pubmed, Embase and the Cochrane Library databases. A total of 36 articles were included, with a sample size of 38,384 bladder cancer patients. Of these, 4,354 patients were reported to have positive surgical margins. Significant associations were detected between positive surgical margins following RC and unfavorable RFS [summary relative risk estimate (SRRE), 1.63; 95% confidence interval (CI), 1.46-1.83;  Bladder cancer is the ninth most commonly diagnosed cancer worldwide and ranks second in terms of incidence among genitourinary malignancies. The American Cancer Society estimated that in the United States there were 76,960 new cases and 16,390 mortalities from bladder cancer in 2016 [Surgical margin status is determined by the presence or absence of tumor tissue in the areas of soft tissue around the surgical specimen. Although the potential associations between surgical margin status and survival outcome have received much attention in the past few years, studies have yielded inconsistent results. Several large retrospective studies have demonstrated that positive surgical margins are an independent predictor of recurrence and eventual mortality from bladder cancer –6. On thIn the present study, a meta-analysis was conducted to summarize the relationship between surgical margin status and bladder cancer outcome after RC based on all published epidemiological studies. The purpose of this meta-analysis was to clarify the association between surgical margin status and survival outcomes of RC, and explore potential sources of heterogeneity across different studies.Figure The main characteristics of the included studies are shown in Table P = 0.105; I2 = 32.1%). The pooled result indicated that the presence of positive surgical margins was associated with poor RFS. The CSS was reported in 26 studies that enrolled a total of 25,804 bladder cancer patients. A random effects model was used due to evidence of heterogeneity among the studies . A significant CSS disadvantage was detected in the positive surgical margin group compared with the negative surgical margin group .In 16 studies, with a total sample size of 10,738 individuals, the associations between positive surgical margin and RFS of bladder cancer patients after RC were reported. A fixed effects model was used, revealing a summary relative risk estimate (SRRE) of 1.63 [95% confidence interval (CI), 1.46-1.83; Figure In sensitivity analyses excluding one study at a time, the SRRE for RFS ranged from 1.59  to 1.71 . Similarly, the SRRE for CSS ranged from 1.77  to 1.86 , and the SRRE for OS ranged from 1.67  to 1.71 . These results indicated that the findings were reliable and robust. No statistical evidence of publication bias was found in this meta-analyses, as assessed by Begg’s and Egger’s tests for RFS , sample size (≥500 vs. < 500), publication year (≥2014 vs. < 2014), duration of follow-up (≥60 months vs. < 60 months), study type (single center vs. multicenter) and geographic region . Although no significant modifiers accounting for the inter-study heterogeneity were detected, the observed heterogeneity in CSS decreased significantly in some models, such as articles published since 2014, multicenter studies, study with sample size >500 cases, and with follow-up duration >60 months. Furthermore, the results in subgroup analyses were consistent with the primary findings , 1.82  and 1.68 , respectively, compared with negative surgical margins. The findings were consistently independent of age, sample size, publication year, follow-up duration, study type and geographical region. The sensitivity analyses indicated that the findings were reliable and robust. In addition, there was no evidence of significant publication bias in these analyses according to Begg’s or Egger’s tests.Heterogeneity is often a major concern in meta-analyses. Significant heterogeneity was detected in the analysis of the CSS model, although there was no evidence of heterogeneity in terms of RFS or OS. Many confounding variables differed across the individual studies, including age, gender, pathological stage and lymph node status; this may explain the observed heterogeneity between the studies. To address the issue of heterogeneity, subgroup analyses were conducted, revealing that the significant variations were reduced in the meta-analysis of article subgroups published since 2014, with a multicenter design, with a sample size >500 patients and with follow-up duration >60 months.Potential risk factors contributing to positive surgical margins are as follows: i) features of advanced cancer, such as lymphovascular invasion, extravesical disease and mixed histology ; ii) surSeveral important strengths of the present study should be noted. Firstly, the meta-analysis included 36 studies with a large sample size to detect more stable associations and provide more reliable results. Secondly, strict accordance with the inclusion and exclusion criteria was maintained, and we also extracted available data from relevant studies that mentioned the relationship between surgical margin status and survival outcomes of bladder cancer patients after RC. Furthermore, the results were found to be reliable and robust through subgroup and sensitivity analyses.However, the study was subject to several limitations. First and foremost, the majority of the included studies were retrospective cohort studies, which made our meta-analysis sensitive to potential confounding variables. Additionally, although the results from the main multivariable model that included the most adjusted confounders were used, there may be residual or unknown confounding variables that were not taken into consideration in the included studies. Secondly, substantial heterogeneity was observed in the meta-analysis of CSS. Although subgroup and meta-regression analyses were conducted to explore the source of heterogeneity, no effect modifier of heterogeneity was found. Thirdly, we were unable to explore the potential differences in associations according to the classification of bladder cancer. It remains unknown whether the findings may vary by tumor subtype or tumor stage, even though some subgroup analyses were conducted. An additional limitation is that the detection methods of surgical margin status were not definitely described in the majority of the included studies. Therefore, a subgroup analysis for detection methods could not be performed. Moreover, other detailed information regarding the features of the margins, such as the location, focality, and microscopic or macroscopic features, were also not presented in the included studies, and could not be further examined.In summary, the present meta-analysis confirms that bladder cancer patients with positive surgical margins, as compared with negative surgical margins, are likely to have poorer RFS, CSS and OS after RC, indicating that surgical margin status may be an independent indicator of the survival outcome of patients with bladder cancer following RC.This meta-analysis was performed in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guidelines . A systeThe eligibility of each study was assessed using the population, intervention, comparator, outcome and study design (PICOS) approach . A studyIn addition to these criteria, only studies that reported survival outcomes, such as RFS, CSS or OS, were considered for inclusion. Case-reports, reviews, expert opinions or meeting abstracts without usable data were excluded. In studies with the same or overlapping population, only the most recent and informative study was included in the meta-analysis.Data from all included studies were independently extracted by two investigators and corroborated by another investigator. The following information was extracted for each included study: first author, year of publication, study type, country, study period, duration of follow-up, sample size, mean age, gender, pathological stage, positive lymph node rate, positive surgical margin rate, and risk estimates of RFS, CSS or OS based on margin status. If one study contained multiple data sets, the one with more adjusted confounders was used. All discrepancies in data extraction were resolved by discussion.2 statistics. P < 0.10 or I2>50% were used to indicate heterogeneity. Random effects models were used for meta-analysis in cases of heterogeneity. Forest plots were also applied to assess the relationships between margin status and survival outcomes of bladder cancer treated by RC.All included studies used multivariable Cox proportional hazards ratio models to examine the associations between surgical margin status and survival outcomes of RC, including RFS, CSS or OS. SRREs and 95% CIs were calculated in order to compare the positive margin group with the negative margin group. Heterogeneity between studies was assessed using the Q and IP < 0.05 was considered to be an indicator of significance, except where specifically noted.Subgroup and meta-regression analyses were performed to examine potential sources of heterogeneity according to age, sample size, publication year, follow-up duration, study type and geographical region. Sensitivity analyses were conducted to assess the robustness of the results by repeating the meta-analysis after omitting one study at a time. Furthermore, funnel plots were inspected for asymmetry, and Begg’s and Egger’s tests were used to assess publication bias. All statistical analyses were conducted using STATA 12.0 software .


Periodontitis results from an inflammatory response caused by accumulative microorganisms in periodontal sites. Several factors are involved in pathogenesis of periodontitis, for example the -889 C/T polymorphism in interleukin-1A gene. This study aimed to evaluate the relationship between this polymorphism and risk of development of chronic periodontitis by a meta-analysis based in new published findings.P <0.05.Thereunto a review in literature was performed in the electronic biomedical and education databases  to studies published before August 2, 2015, the abstracts were evaluated and the data extraction performed by two calibrated examiners. The calculations of the meta-analysis were obtained through statistical software Review Manager version 5.2 with calculation of Odds Ratio (OR), heterogeneity (I²) and Funnel plots with P = 0.0004) with decreased value to heterogeneity . TT genotype was associated to patients with chronic periodontitis . No publication bias was found in this meta-analysis by asymmetry in Funnel plots.In overall, twenty-one case/control studies were selected with 2,174 patients with chronic periodontitis and 1, 756 controls. The meta-analysis showed T allele was associated with chronic periodontitis (OR = 1.22, 95% CI: 1.09, 1.36, This meta-analysis with 2,174 patients with chronic periodontitis and 1, 756 controls evidenced the -889 C/T polymorphism is associated to risk of development of chronic periodontitis with no significant value to heterogeneity to allelic evaluation.Key words:Alleles, odds ratio, periodontal disease, cytokines. Periodontitis is a chronic inflammatory disease caused by accumulative plaque beyond gingival sulcus and host-immune response with involvement of multifactorial process . The disIn physiopathology of chronic periodontitis several inflammatory mediators contribute with damage in periodontal sites as well as connective tissue and alveolar bone loss . For exaIL-1A facilities and amplifies inflammatory response by induction of adhesion molecules to cellular infiltrate  and monoGenetic variations in IL-1A gene were associated with elevated risk to chronic periodontitis by elevated level of this cytokine in gingival fluid , but resThree meta-analysis evaluating a polymorphism -889 C/T) in IL-1A and chronic periodontitis are available in literature ,7,8 and 89 C/T inHowever, since 2013 a significant number of studies have been published -13, thes- Data sourcesA systematic search in literature was performed by three investigators in the electronic biomedical and education databases  to studies published before August 2, 2015 and addressing the association of -889 C/T polymorphisms in IL-1A gene and risk of development of chronic periodontitis. The following combined keywords were used to retrieve the literature: (“interleukin” or “cytokine”) and (“polymorphism” or “-889 C/T” or “rs1800587”) and . No language restriction was placed on the search and all citations of studies were screened to identify additional potential studies.- Inclusion criteriaArticles were included in current meta-analysis if the studies met all the following criteria: [1] Evaluation of the polymorphism cited and risk of development of chronic periodontitis; [2] Studies are case/control design; [3] Genotype frequency documented; [4] Diagnosis of periodontitis confirmed through radiographic findings and clinical evaluation. Studies which did not bring sufficient information about genotype or allelic frequencies or did not respect any point these criteria were excluded.- Data extractionTwo investigators independently reviewed all studies and extracted the data using a standardized form. Data were collected on the authors, year of publication, ethnicity, study design , number of cases and controls, age and subject type in study.- Statistical analysisThe statistical analysis of data was performed with use of Review Manager version 5.2 software  and publication bias with Comprehensive Meta-analysis version 3.3.070 [2014] statistical software.P>0.05) the Fixed-effect model was used to estimate the pooled Odds Ratio (OR). In other hand when heterogeneity was significant  the Random-effects model was used to OR calculation. Both methods the P value <0.05 was considered statistically significant. Begg’s test and Egger’s linear regression test (with P<0.05) were used to evaluate potential publications bias of reported associations with funnel plot asymmetry. All of data in studies were dichotomous data expressed as OR with 95% of confidence intervals (CI) to assess the association between polymorphism in IL-1A gene and periodontitis.The chi-squared based Q statistic test (I²) was used to assess the presence of heterogeneity. When value of I² was not statistically significant  and were outlier in funnel plot graphic, after exclusion, the heterogeneity decreased to be unremarkable , although these studies did not carry out heterogeneity in genotypic evaluation. The forest plots to T allele versus C allele in overall analysis and to C allele versus T allele were shown in figure P = 0.0004) and C allele was associated to control group . In both calculations the Fixed-effect statistical model was used to estimate the pooled Odds Ratio. Besides, TT genotype was associated to patients with chronic periodontitis in overall . The The meta-analysis showed the -889 C/T in IL-1A gene is associated to elevated risk development of chronic periodontitis. In allelic evaluation, four studies ,19,20,23- Sensitivity analysis and Publication biasP = 0.901 and P = 0.791, respectively) as showed by no funnel plot asymmetry in figure To evaluate the individual effect of studies a sensitivity analysis was performed by omitting each study to assess this impact on pooled ORs. No single publication changed the pooled ORs qualitatively, which suggested that results of this meta-analysis were accurate. The Begg’s test and Egger’s linear regression test did not reveal any indication of publication bias in allelic evaluation   .P<0.00001). This finding contradicts others studies carried out in Caucasian populations from Australia  corroborating data previously found in Brazilian population , which showed an OR = 10.9 for individuals that presented previous periodontitis  and C allele was associated to control group , both in overall analysis.In conclusion, this meta-analysis, composed by twenty-one studies in various ethnic groups totaling 2,174 patients with chronic periodontitis and 1,756 controls, showed T allele in -889 C/T was associated to risk of development of chronic periodontitis (OR = 1.22, 95% CI: 1.09, 1.36,


Playing critical roles in immune responses, interleukin-6 (IL-6) has been proposed to be involved in the development of multiple cancers, including prostate cancer. The rs1800795 polymorphism in the promoter of the gene IL-6 can affect the transcription and expression of the gene, becoming a common target in association studies on tumors. We therefore carried out this meta-analysis to further discuss the relationship of this polymorphism with the risk of prostate cancer.IL-6 rs1800795 polymorphism and prostate cancer risk was evaluated using pooled odds ratios (ORs) with their 95% confidence intervals (95% CIs). Q test was adopted to examine between-study heterogeneity, with P < 0.05 as significant level. Subgroup and meta-regression analyses were conducted to explore potential source of heterogeneity. Sensitivity analysis was implemented to test the statistical stability of the final results. In addition, funnel plot and Egger test were employed to inspect publication bias among included studies.Relevant publications were retrieved from the electronic databases. The strength of the correlation between IL-6 rs1800795 polymorphism and prostate cancer risk in total analysis, but a risk-increasing effect of the polymorphism was detected in African-American subgroup under CC versus GG and CC versus GG + GC contrasts  after subgroup analysis by ethnicity.A total of 13 132 cases and 15 282 controls were ultimately incorporated into the present study. Overall estimates revealed no significant relationship between IL-6 rs1800795 polymorphism may enhance the susceptibility to prostate cancer in African-American men. The incidence rate of this cancer varies among regions and races, with about two-third of total prostate cancer cases occurring among men in more developed countries who only account for just 17% of global male population. This malignancy, however, still shows an upward tendency in its incidence in both previously high-risk regions and in relatively low-risk ones during the past few decades. At the moment, the exact etiology of prostate cancer is still poorly understood, but some aspects have been confirmed as risk factors for it, such as age, race, and family history. In addition, the differences of incidence rate between races and regions may be related to lifestyles, environmental conditions, and genetic backgrounds. Existing documents have demonstrated that among material bases for genetic susceptibility to prostate cancer, single-nucleotide polymorphism (SNP) presents a key one.Prostate cancer is 1 of the most commonly diagnosed male malignancies jeopardizing their health, and frequently affects those aged over 60 years.,9 Human IL-6 gene is located at chromosome 7p21–14 with a total length of 5 kb, and contains 4 introns and 5 exons. With the advancement of studies on tumor pathogenesis, the initiation and progression of tumors have been uncovered to be closely related to the loss of regulation of host immune system on tumors. Cytokines play key regulating roles in immune responses, and some of them even directly influence the growth and invasion of tumors. Reportedly, IL-6, as an autocrine or paracrine factor, is able to not only regulate tumor growth via straight effects on tumor cells, but also indirectly promote the growth of tumor cells through affecting host environments, such as inducing antiapoptosis, neovascularization, and acute phase responses.,14 In recent years, the role of IL-6 in the origination and development of malignant tumors has been kept exploring, and abnormal expression of this cytokine has been observed in multiple tumor tissues, such as myeloma, renal cancer, hepatocellular carcinoma, lung cancer, and esophageal carcinoma.Interleukin-6 (IL-6), a bioactive peptide with multiple functions, is mainly originated from mononuclear phagocytes and partly from fibroblasts, T and B lymphocytes, and vascular endothelial cells.IL-6 can affect the transcription of the gene, and thus alter the cytokine production. Therefore, this SNP has been discussed in previous association studies on various cancers. In this meta-analysis, we targeted this polymorphism to examine its role in the susceptibility to prostate cancer.The rs1800795 polymorphism in the promoter of the coding gene 22.1This meta-analysis was conducted in accordance with the checklist of the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines. A systemic literature search was conducted in the databases PubMed, EMBASE, Goggle Scholar, and CNKI using the combination of the following terms: “prostate” or “prostatic,” “cancer” or “carcinoma” or “tumor” or “neoplasm” or “malignancy” or “malignant,” “interleukin-6” or “IL-6” or “ BSF2” or “HGF” or “HSF,” and “polymorphism” or “mutation” or “SNP” or “variant” or “polymorphisms.” To supplement the yield of database searching, we also manually checked the reference lists of relevant papers.2.2IL-6 rs1800795 polymorphism and prostate cancer risk might be qualified for our meta-analysis, and could enter into next circle of eligibility assessment; otherwise, they were excluded from the present study. Predesigned criteria for detailed evaluation contained the following aspects: with a case-control design; stating sufficient information on genotype and/or allele frequencies in cases and controls; and concerning human beings. Articles failing to satisfy any one of the above standards were removed in this course. As for reports with overlapping data, the one would be finally selected which covered the most comprehensive information.Studies focused on the relationship between 2.3P value for Hardy–Weinberg equilibrium (HWE) in controls.Two reviewers took charge of data extraction from all eligible studies, and completed cross-check on the recorded data to guarantee their accuracy. If disagreements occurred in this process, they would be settled through discussion between these 2 reviewers until consensus was achieved on each item. Many aspects of included studies were extracted, including first author's name, year of publication, country of origin, ethnic line, control source, genotyping method, genotype and/or allele frequencies in case and control groups, and 2.4IL-6 rs1800795 polymorphism and the susceptibility to prostate cancer was appraised through calculating pooled odds ratios (ORs) with the corresponding 95% confidence intervals (95% CIs). Heterogeneity between selected studies was inspected using chi-square-based Q test, with P value more or less than 0.05, representing the absence or presence of significant heterogeneity. When significant heterogeneity existed, we would implement meta-regression analysis to identify its possible source. When P > 0.05 in Q test, fixed-effects model was chosen for calculating summary ORs; or else, random-effects model was applied. Moreover, subgroup analyses, which were stratified according to the patients’ ethnicity , and source of controls  were performed to explore potential sources of heterogeneity and the differences among them. In addition, sensitivity analysis was conducted through sequential deleting each of included studies so as to verify the stability of overall estimates. Publication bias across enrolled studies was investigated with both Begg funnel plot and Egger regression test.STATA 12.0 software  was utilized to complete all data syntheses in this meta-analysis. The intensity of the relationship between 33.1–34 enrolling 13,132 cases and 15,282 controls. Table In the initial search, 195 potentially relevant publications were identified, and then 13 duplicates were excluded from the current meta-analysis  manifested that such heterogeneity could be mainly attributed to the study by Mandal et al.Under the other 1 comparison, the fixed-effects model was chosen for OR calculation in view of the absence of significant heterogeneity.3.4We recalculated summary ORs after expunging each of the eligible studies in turn, and then compared them with original overall estimates. The comparison detected no qualitative alteration between the effects, verifying that our results were statistical robust and stable.3.5P = 0.879). Therefore, publication bias across the selected studies in this meta-analysis was negligible.In visual inspection of Begg funnel plots, we found the shape of these plots seemed symmetrical Fig. . Moreove4 Previous studies have shown that the IL-6 polymorphism (rs1800795) may predispose to prostate cancer and influence disease severity.,29 The rs1800795 polymorphism located in the promoter of the coding gene IL-6, has been reported to affect its gene transcription and thus change serum IL-6 levels.Up to now, the etiology of prostate cancer is still unclear, and relevant researches have put forward multiple potentially relevant aspects, involving environmental factors and genetic factors. The polymorphism, as inherent mechanisms of human diseases, has been used to explain the differences in incidence, clinical manifestations, and response to treatment of tumor patients, including prostate cancer. As a pleiotropic cytokine, IL-6 can regulate many cell functions, including immune defensive mechanism, cell proliferation, and differentiation, and also the production of haemocytes. In addition, it also has close relationship with the origination and progression of multiple tumors, affecting the development of tumors through influencing cells’ ability of adhesion and activity, the formation of thrombi, the expression of tumor-specific antigens, and the proliferation of tumor cells.–38 Several studies have shown an association between IL-6 gene polymorphisms and the risk of prostate cancer, but the results are inconclusive. For example, Mandal et al, in their study, found different effects of this polymorphism on the risk of prostate cancer. Specifically, the GG genotype of the polymorphism increased the risk in Caucasian subjects, whereas the CC genotype displayed a similar trend in African-American cases. In other words, IL-6 -174G>C polymorphism might play totally opposite effects in different races.Evidences exhibit that IL-6 plays an important role in the transformation of prostate cancer from hormone-dependent to hormone-independent, thus attenuating the efficiency of endocrine therapy.IL-6 rs1800795 polymorphism with the susceptibility to prostate cancer in total analysis. However, the polymorphism significantly elevated the risk in African-American group after stratification analysis by ethnicity under CC versus GG and CC versus GG + GC contrasts, which was consistent with the findings from the study by Mandal et al. In other subgroups, no significant relationship was detected. Generally, the findings from this meta-analysis were relatively reliable because they not only were obtained on the basis of a larger sample size of 13,132 cases and 15,282 controls, but also were tested through some examinations. However, significant heterogeneity was detected among included studies, and the results of meta-regression analysis manifested that such heterogeneity could be mainly attributed to the study by Mandal et al. So the random-effects model was engaged in calculating pooled ORs under these cases. Moreover, sensitivity analysis was conducted by expunging each of the eligible studies in turn to see whether a particular omission could influence the overall estimates. The overall estimates expressed no qualitative change after the deletion of the study by Mandal et al, and similar results were observed after removal of other included studies during the sensitivity analysis, confirming that our results were statistical robust and stable. As for possible reasons for the study by Mandal et al presenting the source of significant heterogeneity, we speculated that their finding about the opposite effects of IL-6 rs1800795 polymorphism on prostate cancer risk in 2 different descent groups itself might partly contribute to the occurrence of the heterogeneity. In the present meta-analysis, limited number of patients in various ethnicity groups and different sources of controls may contribute to relatively extended CIs. Future studies that include a larger number of patients with better study designs need to be conducted to clarify this important issue. When it came to publication bias, neither funnel plots nor statistical data from Egger test supported the existence of significant bias. Based on these tests, we believed our conclusion had certain strength.To statistically discuss this relationship, we designed the present meta-analysis based on previously published studies on this topic. After data syntheses, we found no significant correlation of That having been said, the results from the current meta-analysis still should be applied warily due to some inevitable limitations in our study. First of all, we only selected eligible articles from those previously published, so some relevant papers unpublished might be missed, generating certain publication bias, though not detected even with funnel plot or Egger test. Apart from this, detailed subgroup analyses were not performed according to important factors involved in the risk of prostate cancer, such as age, smoking status, alcohol consumption, and family history, owing to restricted information in original articles. Hence, the final effects from this meta-analysis might have some bias to a certain degree. Furthermore, plausible effects of gene–gene and gene–environment interactions on the cancer risk were not taken into account because of limited information as well.IL-6 rs1800795 polymorphism may not have an independent influence on the susceptibility to prostate cancer in general population, but it may significantly increase the risk in African-American males. In view of the above mentioned restrictions, these results need to be further verified by studies fully considering the effects of major factors involved in the cancer incidence.In summary,


PTEN) gene. We searched the PubMed, EMBASE, Cochrane library and the national knowledge infrastructure of China (CNKI) databases and identified 18 eligible case-control studies with 5458 cases and 6003 controls for rs701848 as well as 5490 cases and 6209 controls for rs2735343. Our analyses demonstrated that cancer risk was associated with rs701848 in the recessive model  and with rs2735343 in the dominant model . Subgroup analysis showed that in Asian subjects, carrying the C allele of rs701848 or GG genotype of rs2735343 was associated with increased cancer risk. Moreover, Asian subjects carrying the TC/CC genotype or C allele of rs701848 were associated with increased risk of esophageal squamous cell cancer. This meta-analysis indicates that the PTEN rs701848 (CC) and rs2735343 (GG) polymorphisms are associated with increased cancer risk in Asian subjects.We performed this meta-analysis to analyze the cancer risk to individuals carrying the rs701848 and rs2735343 single nucleotide polymorphisms (SNPs) in the phosphatase and tensin homolog ( PTEN) is also known as mutated in multiple advanced cancers 1 (MMAC1) or TGF-β regulated and epithelial cell-enriched phosphatase 1 (TEP1) and is a tumor suppressor gene [Phosphatase and tensin homolog  are the most common type of genetic variations that involve change in a single nucleotide in a gene or associated genetic elements, which affect gene expression . A numbeblastoma –19.PTEN SNPs, rs701848 and rs2735343 is controversial. The rs701848 SNP is associated with increased risk of breast cancer (BC) [PTEN SNPs, rs701848 and rs2735343.The relationship between cancer risk and two cer (BC) , renal ccer (BC) , coloreccer (BC) , and esocer (BC) . Howevercer (BC)  and hepacer (BC) . Moreovecer (BC) . Subjectcer (BC) . HoweverPTEN SNPs. We then assessed the remaining 53 reports in greater detail and excluded 35 articles that did not satisfy the eligibility criteria. Finally, 18 eligible case-control studies were included in our meta-analysis [We searched the PubMed, EMBASE, Cochrane library and the national knowledge infrastructure of China (CNKI) databases and identified 1230 articles. After removing the duplicate articles, 892 articles still remained for further evaluation. Then, we reviewed article titles and abstracts and excluded 839 reports that were not related to cancer risk and analysis –37  for rs701848 polymorphism in recessive model was analyzed by fixed-effects model. Since data was heterogeneous, random effects model was used to analyze the significance of pooled OR for rs701848 in homozygous, heterozygous, dominant, and recessive models and in all models for rs2735343 , and sample size, respectively. We observed that Asian individuals with rs701848 C allele were associated with 1.105-fold increased cancer risk than the non-Asian population , indicating that there exists genetic heterogeneity between different cancer types.We conducted meta-regression analysis based on ethnicity, cancer type, source of controls, genotyping methods, quality score (at the median cut-off point of 8), and sample size parameters to determine the factors that are critical for association of the We performed Egger's Table  and BeggPTEN have been reported as prognostic factors in several cancers [PTEN gene are associated with increased risk of digestive tract cancer [PTEN SNPs also play important roles in tumorigenesis. The PTEN rs11202586 SNP is associated with increased risk of testicular germ cell tumor [et al showed that the PTEN rs3830675 SNP was associated with colorectal cancer in patients that consumed alcohol and smoked [PTEN exerts its tumor suppressor function by acting as a negative regulator of the mTOR/Akt signaling pathway . Mutatio cancers , 39, 40.t cancer . PTEN SNll tumor . Han et d smoked . In thisPTEN gene and increase cancer risk by probably influencing splicing, protein expression and cell cycle [PTEN mRNA stability [PTEN were known to participate in tumorigenesis, their relationship with cancer risk was unknown [et al [et al [PTEN rs701848 SNP contributed to ESCC risk, especially, the individuals carrying CC genotype in PTEN rs701848 have a 1.813-fold increased cancer risk of ESCC. Our conclusion was different from Ma's study that investigated 206 ESCC cases and controls each and concluded that the rs701848 CC genotype was not associated with ESCC risk [Both rs701848 and rs2735343 SNPs are located in the intron and non-coding region of ll cycle . The rs7tability . Althoug unknown , 46. Jann [et al  and Xu el [et al  showed tSCC risk . We perfSCC risk , 33, 35.SCC risk  and Ma'sSCC risk  findingsCancer is a genetic disease because the underlying causes include somatic mutations, chromosome translocations, gene amplification, and epigenetic changes –49. A siThe major drawback of our meta-analysis was that it limited to individuals of Asian descent. Therefore, the effects of rs701848 and rs2735343 on non-Asian populations need to be studied in well-designed and large scale case-control studies. In conclusion, we demonstrate that the C allele of rs701848 and G allele of rs2735343 in PTEN gene increases cancer risk in Asian populations.Observational Studies in Epidemiology (MOOSE) group [PTEN rs701848 or rs2735343 with cancer risk.This meta-analysis was performed according to the protocols of the E) group . Two resThe inclusion criteria for eligible articles were as follows: (1) the articles assessed the association between rs701848 or rs2735343 and cancer risk; (2) it was a case-control study; (3) study subjects diagnosed with malignant tumors were histologically confirmed; (4) sufficient data was available to calculate OR and the corresponding 95% CI. When the data in the articles was insufficient, we attempted to obtain the missing data from the first or corresponding authors via email.As mentioned above, two reviewers, SDD and LJH independently searched articles, extracted data and assessed the quality. If there was a controversy, a third researcher, ZQ, was involved to resolve the issue by discussion. The extracted data included first author, published year, country of origin, ethnicity, cancer type, genotyping method, characteristics of cases and controls, source of controls, and P value for HWE. The quality of studies was assessed by the Newcastle-Ottawa quality assessment scale for observational studies. The assessment scale had three categories, namely, selection, comparability, and exposure, which altogether contained eight items. A study was awarded a maximum of one point for each parameter within the selection and exposure categories. A maximum of two points were awarded for comparability. The maximum obtainable score was nine.I2 test. A I2 > 50% suggested heterogeneity [P< 0.05 for Begg's and Egger's tests indicated significant publication bias. Data were analyzed using STATA 12.0  and Review Manager 5.3  software.The control group was analyzed by chi-square test and P > 0.05 was in accordance to HWE . The relogeneity . A randoogeneity . The Taqogeneity . Publicaogeneity  and Eggeogeneity  tests, r


ATXN2 gene in amyotrophic lateral sclerosis (ALS). Two new case-control studies, a British dataset of 1474 ALS cases and 567 controls, and a Dutch dataset of 1328 ALS cases and 691 controls were analyzed. In addition, to increase power, we systematically searched PubMed for case-control studies published after 1 August 2010 that investigated the association between ATXN2 intermediate repeats and ALS. We conducted a meta-analysis of the new and existing studies for the relative risks of ATXN2 intermediate repeat alleles of between 24 and 34 CAG trinucleotide repeats and ALS. There was an overall increased risk of ALS for those carrying intermediate sized trinucleotide repeat alleles , with an exponential relationship between repeat length and ALS risk for alleles of 29–32 repeats . No relationship was seen for repeat length and age of onset or survival. In contrast to trinucleotide repeat diseases, intermediate ATXN2 trinucleotide repeat expansion in ALS does not predict age of onset but does predict disease risk.We investigated a CAG trinucleotide repeat expansion in the  All trinucleotide repeat diseases show neurological features and include Huntington's disease , a neurodegenerative disease of upper and lower motor neurons, but this association is only seen for repeats of intermediate size, below the range usually associated with spinocerebellar ataxia (34 repeats or more) but above the normal range  were collected from a consecutive clinical case series obtained from King's College Hospital (n = 116), from the Motor Neurone Disease Association DNA Biobank (n = 1051), and from Queen Mary University of London and UCL Institute of Neurology (n = 307). All patients were diagnosed as having definite or probable ALS according to the El Escorial criteria. The DNA samples of 567 neurologically normal controls, matched to patients for gender, age, and geographical region, were obtained from the MRC London Neurodegenerative Diseases Brain Bank, the Institute of Psychiatry, Psychology and Neuroscience (n = 68), from the National Institute for Health Research Mental Health Biomedical Research Centre and the Dementia Unit at South London and Maudsley NHS Foundation Trust and the Institute of Psychiatry, King's College London (n = 306), and from the Motor Neurone Disease Association (n = 193).ATXN2 repeat size.A second case-control population-based set was obtained in collaboration with the University Medical Center Utrecht, the Netherlands, with a total of 1328 unrelated ALS cases (23 with a family history in a first degree relative) and 691 neurologically normal controls, matched to patients for gender, age, and geographical region . Samples2.2Informed consent was obtained from all included in the study. The study was approved by the Trent Research Ethics Committee 08/HO405/60 and by the Medical Ethics Review Board at the University Medical Center Utrecht 05_067/E.2.3ATXN2 CAG trinucleotide repeat region was amplified according to a previously published PCR protocol . The new case-control studies from UK and Dutch populations were included in the analysis. Bias in individual studies was evaluated using the Newcastle-Ottawa Scale questionnaire for Quality Assessment of Nonrandomized Studies  and Cochrane's Q-test of heterogeneity. I2 > 50% or p < 0.05 for the Q-test were taken as indicative of significant heterogeneity. We used a fixed effects model following the assumption that all studies had a common genetic effect and that specific findings of each study were due to random sampling. However, to control for any possible difference across studies, the fixed effect model RR estimates at each threshold were compared with RR estimates assessed using a random effects model. A sensitivity analysis, leaving out one study at a time, was performed to test the robustness of the meta-analysis and assess the influence of individual studies on the overall result for each allele. Possible sources of heterogeneity across studies were explored using subgroup analysis using source of the control group (population based vs. nonpopulation based) and geographic location  as covariates. Meta-regression was used to further investigate differences between population-based and nonpopulation-based subgroups. Funnel plots were generated for each intermediate repeat allele to analyze the intervention effect from individual studies against study size. A resulting p < 0.05 was considered as indicative of the presence of small-study effects. Correlation between age at onset and the CAG trinucleotide repeat length of risk alleles was tested in the new British and Dutch ALS cases, and in published data sets for which age at onset data were available. Two-tailed Fisher exact tests were used to test for differences in demographic and clinical characteristics of patients by ATXN2 repeat size. ANOVA was used to compare ages at onset for different repeat sizes, as well as by using SNPs rs695871 and rs695872, previously shown to associate with age of onset. Kaplan-Meier survival analysis and a log-rank test were used to compare survival time between groups. We compared the fit of an exponential model with the fit of a linear model using Akaike information criteria and Bayesian information criteria. Comparison of model values for either measure can be used to assess fit provided the values differ by more than 10. The model with the larger value has less support (http://www.R-project.org) was used to test the hypothesis that the relationship between ATXN2 CAG repeat length and ALS risk fitted an exponential model.Relative risks (RRs) were approximated by the odds ratio, generated with corresponding 95% confidence intervals (CIs), by meta-analysis using a Cochran-Mantel-Haenszel chi-square test comparing the case-control counts for a specific allele with the pooled counts for alleles of 23 repeats or fewer across the different published and unpublished studies. RR was then estimated comparing pooled counts of risk alleles with counts of alleles of 23 repeats or fewer. Where a cell contained zero observations, a continuity correction of 0.5 was applied. The sample size was considered adequate to measure the effect size since each individual study in the meta-analysis measured the effect, and the addition of further samples would increase power further. We assumed that all studies were estimating the same common effect and estimates varied only because of chance differences in sampling patients. To assess our assumption, heterogeneity between studies was estimated using the I support . Meta-an3ATXN2 trinucleotide CAG repeats were analyzed in 1474 ALS cases and 567 neurologically normal controls from the UK and in 1328 ALS cases and 691 neurologically normal controls from the Netherlands. The distribution of allele frequencies is shown in ATXN2 intermediate expansions and C9orf72 expansion. Six patients  had intermediate ATXN2 expansion and pathological expansion of C9orf72. Exclusion of these patients from analyses did not change the overall findings.All cases were tested for the copresence of ATXN2 repeat expansion in ALS, contacting authors for raw data where necessary, and including studies for analysis based on strict criteria  to an overall I2 value of 0% (heterogeneity p = 0.51) but did not influence the overall RR estimate. No study exclusion could explain the significant heterogeneity present in the analysis of the 24 repeat allele. No other studies had any effect on heterogeneity or RR estimates, including 2 studies that used young controls below the age of risk . Nor were there any associations when SNPs rs695871 and rs695872, previously shown to associate with the age of onset, were tested.Next, we tested the relationship between repeat length and age of ALS onset in the 4 different populations for which data were available. In keeping with previous findings, and in contrast to trinucleotide repeat diseases, we found no evidence for such a relationship  or Dutch  cohorts.We also assessed differences in demographic and clinical characteristics between patients with CAG repeats <29 and patients with CAG repeats ≥29 in both British and Dutch cohorts. No significant difference was found in gender, age at onset, or site of onset . No sign4ATXN2 trinucleotide repeat alleles in ALS is 29–33. An unexpected and important finding is that the risk of ALS increases exponentially with allele repeat size until the border with spinocerebellar ataxia risk, even though the age of onset does not change. This may appear surprising but is entirely consistent with current hypotheses of ALS causation, in which the odds ratio conferred by genetic variants and age of onset of first symptoms are not correlated, even within the same family , and has implications for genetic counseling of individuals carrying intermediate size repeats. Two published meta-analyses, one of 12 and another of 13 studies, reported significant association with ALS for pooled analyses of  repeats . Anothercontrols .ATXN2 repeat length in ALS have not shown a relationship with age of onset (ATXN2 intermediate length repeats on ALS survival for those with 31 or more repeats (It has been suggested that shorter intermediate repeats may be protective for ALS . Our datof onset . In this repeats . We did ATXN2 trinucleotide repeat by direct sequencing but become less likely as the length of the repeat increases. When CAA interruptions occur in repeats of size 33–40, the phenotype is usually of a dopa-responsive Parkinsonism (The mechanism by which expanded trinucleotide repeats cause disease does not appear to be the same for all such diseases, even though all trinucleotide repeat expansions result in neurological dysfunction. In some cases, there is a loss of function, for example through hypermethylation , whereasinsonism . One mecinsonism . Thus th5ATXN2 gene in the size range exclusively for ALS risk represents an exponentially increasing risk for each additional repeat.Our study increases the breadth of known effects of trinucleotide repeat expansion size, adding disease risk to the existing correlations with age of onset and disease severity. Thus, the main finding presented here is that trinucleotide repeat expansion in the The authors have no conflicts of interest to disclose.


To meta-analyze complication rate in computed tomography (CT)-guided transthoracic lung biopsy and associated risk factors.Four databases were searched from 1/2000 to 8/2015 for studies reporting complications in CT-guided lung biopsy. Overall and major complication rates were pooled and compared between core biopsy and fine needle aspiration (FNA) using the random-effects model. Risk factors for complications in core biopsy and FNA were identified in meta-regression analysis.p < 0.001). For FNA, larger needle diameter was a risk factor for overall complications, and increased traversed lung parenchyma and smaller lesion size were risk factors for major complications. For core biopsy, no significant risk factors were identified.For core biopsy, 32 articles  were included and for FNA, 17 . Pooled overall complication rates for core biopsy and FNA were 38.8 % (95 % CI: 34.3–43.5 %) and 24.0 % (95 % CI: 18.2–30.8 %), respectively. Major complication rates were 5.7 % (95 % CI: 4.4–7.4 %) and 4.4 % (95 % CI: 2.7–7.0 %), respectively. Overall complication rate was higher for core biopsy compared to FNA  contains supplementary material, which is available to authorized users. In the US, lung cancer screening by low-dose computed tomography (CT) is recommended for people at high risk , and theCT-guided transthoracic lung biopsy is a widely accepted procedure , 4, althComplication rate and diagnostic performance are the two main factors in choosing a diagnostic procedure. To determine the role of CT-guided transthoracic lung biopsy in the work-up of screen-detected lung nodules, it is imperative to know how safe the procedure is. We conducted this meta-analysis to 1) determine the complication rate, and 2) identify risk factors for complications of CT-guided core biopsy and FNA.This study was conducted according to the PRISMA guidelines for systematic reviews and meta-analyses .A literature research was performed from January 2000 to August 21, 2015, on PubMed, Embase, Web of Science, and the Cochrane Library using variations of the combination of the following search terms: (biopsy OR FNA) AND (transthoracic OR CT-guided) AND (lung cancer) AND CT. Please see e-Table Inclusion criteria were: (a) reporting of complications of at least 50 procedures; (b) differentiation in complications between core biopsy and FNA if both techniques were used; (c) the study was not a subset of patients from other included studies; (d) adequate complication monitoring. Complication monitoring was considered adequate if directly following the procedure a CT scan was acquired, plus a CT scan or chest radiograph 2 to 4 h after the procedure. Studies were excluded during screening if they clearly addressed a different topic, were case reports, conference abstracts, reviews or editorials, or if they were not published in English.A standardized extraction form was used to collect the characteristics of the study regarding patients, nodules, procedures and complications, and how complication monitoring was performed. Two authors  independently extracted these data, with disagreement resolved by consensus. The database was split according to the biopsy method, and all analyses were performed separately for core biopsy and FNA.The methodological quality of the studies was assessed using the Newcastle-Ottawa Scale for nonrandomized studies, with results in a score of 0 to 9 . Two autComplications were classified as minor or major according to the Society of Interventional Radiology (SIR) Guidelines . Minor cStudy-specific risk factors for overall complications and for major complications that were examined are listed in Table I2 index [I2 values of 0, 25, 50, and 75 % were defined as no, low, moderate, and high heterogeneity, respectively. To identify sources of heterogeneity, the effect of potential risk factors on between-study heterogeneity was investigated. The R2 equivalent was determined to express the true variance explained by the model, as a proportion of the total true variance. The risk of publication bias was visually assessed with funnel plots of major complication rate against the sample size of the individual studies.Heterogeneity in the overall complication rate between studies was tested and quantified using the 2 index . I2 valp > 0.10 are presented with odds ratio (OR) and 95 % confidence interval (95 % CI). Statistical significance was set at p < 0.05. All statistical analyses were conducted with Comprehensive Meta-Analysis .Pooled complication rates for core biopsy and FNA were calculated using the random-effects model, weighted by the inverse variance of the studies, under the assumption of heterogeneity. Across subgroups, a common among-study variance component was assumed, and subgroups were combined using a fixed effect model. Forest plots were made for major complications and for all types of complications separately. Differences in complication rate between core biopsy and FNA procedures were assessed using regression analysis. All recorded study, patient, nodule, and procedural characteristics as listed in Table p = 0.917). The case control studies (n = 27) scored least points for representativeness of cases (19/27) and for comparability of cases and controls (20/54). The cohort studies (n = 18) scored least points for comparability of cohorts (26/36). Tables Figure horts 26/6. Tablesrols 20/5. The coh2 = 93.74 % = 38, p < 0.001) and for FNA I2 = 95.3 % =17, p < 0.001). For core biopsy no sources of heterogeneity were identified. For FNA, needle diameter and nodule size were sources of heterogeneity, explaining 17 and 22 % of between-study variance, respectively. The funnel plots are shown in Figs. The heterogeneity between studies was high; for core biopsy Ip < 0.001). Respective rates for major complications were 5.7 % (95 % CI: 4.4–7.4 %) and 4.4 % (95 % CI: 2.7–7.0 %) (p=n.s.). In total, only five studies [Table  studies , 44, 50 p-value < 0.20. For FNA procedures, larger needle diameter was a risk factor for overall complications, with an odds ratio of 0.70  per gauge. When analyzing the needle diameter categorically, an FNA needle gauge of 22 or higher resulted in decreased odds of overall complications of 0.30  compared to lower needle gauges. Figure p = 0.017), and increased traversed lung parenchyma increased the risk of major complication . The meta-analysis of core biopsies did not reveal significant risk factors for overall complications or for major complications.Table This meta-analysis determined the complication rate of CT-guided core biopsy and FNA procedures and identified risk factors for complications. For CT-guided core biopsy the pooled rate of pneumothorax was 25.3 %, of pneumothorax requiring intervention 5.6 %, of pulmonary haemorrhage 18.0 %, and of hemoptysis 4.1 %. For FNA procedures these rates were lower, 18.8, 4.3, 6.4, and 1.7 %, respectively. This difference was significant for all minor complications. For FNA, larger needle diameter, smaller lesion size, and increased traversed lung parenchyma were risk factors for complications.Quality Improvement Guidelines for Percutaneous Needle Biopsy, the SIR and ACR published an estimated pneumothorax rate of 12–45 % and a chest tube placement rate of 2–15 % [To our knowledge, no prior meta-analysis has studied the complication rate of CT-guided lung biopsy. Two large studies based on multi-centre procedures have been published. Wiener et al.  estimatef 2–15 % . Again, Hemothorax rate could not be estimated reliably, because it was reported in only six studies. Other rare major complications such as needle tract seeding, air embolism and death were not reported by any of the included studies. Non-included studies reported a range of 0.02–0.4 % for air embolism , 57, 0.0Comparing core biopsy with FNA is not straightforward. Core biopsy and FNA have their own advantages when used to diagnose lung lesions. Still, overall, complications occurred less often in FNA procedures . For major complications this correlation was not significant, although a similar trend was visible in favour of FNA. Included studies that compared FNA with core biopsy , 60 did In this study, a smaller lesion size and an increased distance traversed through lung parenchyma were found as risk factors for major complications in case of FNA. Patient and nodule characteristics most often mentioned as risk factors are older age, presence of emphysema, smaller lesion size, increased lesion depth, non-pleural contact, and smaller pleural-needle angle , 61–63. In contrast, a meta-analysis can find risk factors in study-specific characteristics such as needle size, use of coaxial needle, number of biopsies, use of CT-fluoroscopy, on-site cytology, number of operators, operator experience, and institute frequency—factors that would be hard to identify in a single-cohort/institute, retrospective study. There was no difference in risk of complication between high and low volume centres. Also, no significant correlation between the sample size, number of biopsies, number of operators, operator experience, use of coaxial needle, CT-fluoroscopy, on-site cytology, or biopsy site down technique and (major) complication rate was found.p = 0.171). CT-fluoroscopy is generally reported to have a lower complication rate due to shorter procedure time and fewer needle passes [Only papers using CT guidance  were included in this study. Conventional CT guidance offers the advantage of a simulated 3D view making it easy to look along the needle path. Also, there is no ionizing radiation exposure for the operator. CT-fluoroscopy offers the advantages of a near real-time imaging feedback as the needle is being inserted. It is, however, associated with an increased patient and operator dose , 64, 65.e passes , 33, 64.n = 10; FNA: n = 3). Because of the low number of studies reporting operator experience, and because only overall mean operator experience was reported (so not per operator), it was not possible to further study this potential risk factor in our meta-analysis.Intuitively, operator experience is thought to influence complication rate. In a large single-cohort study, Yeow et al.  reportedp = 0.164), and for FNA , but none of these correlations were significant.The use of coaxial needles has the advantage of decreasing the number of pleural passes. However, it results in a prolonged connection with the pleura which might lead to increased parenchymal damage due to respiratory motion. Also, it increases the outer needle diameter. None of the included studies specifically investigated the effect of coaxial needles on complication rate. Two other studies did , 67, butrapid needle-out patient-rollover. Kim et al. [n = 147) with the biopsy site downwards during the procedure, using a special table, after which they stayed in biopsy site down position for approximately 15 min. They report a considerable drop in pneumothorax rate compared to the standard procedure, which also suggests that patient (re) positioning in the initial minutes after, or even during, biopsy is critical.The biopsy site down technique has been cause for some debate; although some papers report no difference in complication rate when repositioning the patient after the procedure , others m et al.  recentlym et al.  have beefine needle aspiration. When categorized according to this definition, the use of fine needles compared to larger needles decreased the risk of complications by 70 % . For major complications, however, needle size was not a significant risk factor. For core biopsy procedures no significant risk factors were found.For overall complications in FNA, the use of larger needles was a risk factor. Per increased needle gauge, the risk decreased by 30 % . According to the guidelines of the SIR  only proStudies were only included in this meta-analysis if they reported adequate monitoring of complications. Although that resulted in the exclusion of 75 studies, it made sure that no studies were included that underreported their complication rate. Because chest radiography has demonstrated to miss a significant number of pneumothorax cases after CT-guided lung biopsy compared to CT , an initWe investigated potential sources of heterogeneity, but much of the variance between studies could not be explained. Therefore, a random-effects model was used to pool complication rates. This makes the estimates more reliable as it favours larger studies relatively less, compared to a fixed effects model. Also, outliers do not get weighted as heavily as they otherwise would. Overall, we expect that the provided pooled complication rates are accurate estimates of actual complication rates.This study has some limitations. It was not designed to compare specifically the complication rates of core biopsy with FNA. Most included studies only report complications of one method, so usually no controls in the same population are available. Also, although the quality of the studies included in both groups is generally high according to the NOS score, there are potential sources of bias within the studies. In the past, FNA would be preferred over core biopsy for small nodules, because core samples of high quality were considered hard to obtain. It has previously been shown that smaller lesions are more likely to result in complications, which might cause the pooled complication rates of core biopsy and FNA to be biased , 40. HowPapers were only included if complications of FNA and core biopsy were presented separately. Sometimes both techniques are used in the same setting in an effort to increase the diagnostic performance, which can inadvertently lead to a higher rate of complications –72. ThisLastly, this study could not determine the rate of complications that occur infrequently or require a long term follow-up of the patient, such as death, air embolism and needle tract seeding.In the specific context of diagnostic work-up of lung nodules detected in CT-screening FNA should be favoured over core biopsy. This is especially the case for 22-gauge needles, with which the risk of complications decreases greatly. Also, studies have shown that diagnostic yield does not decrease when using smaller FNA needles , and advIt should be considered that a smaller lesion size is a risk factor for major complications, and that nodules detected by lung cancer CT-screening, needing work-up, are generally smaller in size . HoweverIn order to compare the complications of core biopsy and FNA properly, only randomized controlled trials or even only prospective studies comparing both techniques should be included in the meta-analysis. However, randomized controlled trials comparing these techniques have not been published, and only two prospective studies compared complication rates. Therefore, well-designed randomized controlled trials would be recommended to definitively compare the safety of CT-guided lung core biopsy and FNA.For CT-guided lung biopsy the overall complication rate is acceptable and the major complication rate is low. Minor complications occur more often with core biopsy compared to FNA. For major complications this difference is not significant. In cases of FNA, larger needle size is a risk factor for overall complications, and risk factors for major complications are smaller lesion size and increased traversed lung parenchyma. CT-guided lung biopsy, and particularly FNA with small needles, can be an important diagnostic tool, with a low major complication rate.ESM 1(DOCX 119 kb)Below is the link to the electronic supplementary material.


A number of studies have investigated the association between androgenic alopecia (AGA) and cancer risk, but they have yielded inconsistent results. Therefore, this study was conducted to explore this controversial subject.A literature database search was performed according to predefined criteria. An odds ratio (OR) or a hazard ratio (HR) with 95% confidence intervals (CIs) was retained to evaluate the relationship between the incidence of cancer or cancer-specific mortality and categories of AGA. Then a pooled OR or HR was derived.The pooled results showed that no specific degree of baldness had an influence on the incidence of cancer or cancer-specific mortality. However, AGA, especially frontal baldness, with the incidence of testicular germ cell tumor (TGCT) . A significant increase of risk was observed in relation to high grade prostate cancer (PC)  and vertex with/without frontal baldness was associated with PC risk.The study results supported the hypothesis that AGA is negatively associated with TGCT risk and suggested an overlapping pathophysiological mechanism between them, while the viewpoint that AGA can be used as a phenotypic marker for PC risk was poorly supported.The online version of this article (10.1186/s12885-018-4194-z) contains supplementary material, which is available to authorized users. Androgenic alopecia (AGA) is characterized by nonscarring progressive reduction in the diameter, pigmentation, density, and length of hair from frontotemporal and/or vertex regions of the scalp in a distinctive pattern . It is aA high prevalence of abdominal obesity, hypertension, and lower high-density lipoprotein was found in patients affected by AGA . MoreoveFor the past two decades, a considerable number of publications have studied the relationship between AGA and hormone-related cancer, but they haven’t yielded consistent results. Amoretti et al.  conducteandrogenic alopecia, alopecia, baldness, bald or balding in combination with cancer, tumor, or neoplasm in combination with risk, incidence, or mortality. No language or country filters were imposed. The details of the searching terms were listed in  independently retrieved the information from all eligible records. The methodological quality was assessed by two authors (W.L and L.S) using the Newcastle-Ottawa Scale (NOS) with 0–3 scores defined as low quality, 4–6 scores as moderate quality, and 7–9 scores as high quality. For each study, the participant and study characteristics, number of subjects, type of controls, study design, follow-up time, type of cancer, means of AGA assessment, method of case confirmation, cohort/control selection, and AGA categories were extracted and transformed into the specially designed forms. Disagreements or uncertainties were resolved by the reviewers’ re-verification of the data. If an agreement was not still reached, an additional adjudicator (S.D) was invited into the discussion.If available, multivariate-adjusted risk estimates were used for each study; otherwise, unadjusted RRs, which were calculated according to exposure distributions given in the papers, were utilized. When a study only provided risk estimates from the comparison of a subset, i.e. a specific category of hair pattern such as frontal/vertex baldness with the group of no baldness, the alternative estimates were synthesized as the summarized estimates for overall exposure. If risk estimates were presented as RRs or ORs, combined estimates were generated using the method proposed by Hamling et al. , and if 2-based Q statistics and I2 for statistical significance of heterogeneity, pooled ORs and HRs with 95% CI were calculated using a fixed-effect model (Mantel-Haenszel method) or random-effect model (DerSimonian-Laird method). A P value of Q statistics > 0.10 and I2 < 50% indicated little heterogeneity. Sensitivity analysis was performed to assess the influence of a single trial in the meta-analysis estimated by sequential omission of individual trials. Publication bias was assessed with a funnel plot and the Egger regression. An overall meta-analysis was carried out in all included studies, and then a specific type of cancer with over two articles was included as the subtype that was performed for further meta-analysis. Prespecified subgroup analyses were performed by, reference age  for baldness assessment, pattern of baldness , amount of baldness [1st stage: I; 2nd stage: II, IIa; 3rd -4th stage: III, IIIa, III-vertex, IV, Iva; 5th–7th stage: V, Va, VI, VII which were measured by Hamilton-Norwood scale  was used for statistical analysis.A two-sided The literature retrieval identified 1562 records in Pubmed and 4476 in Embase. Ultimately, 20 publications –35 were In total, six cohort studies , 24, 27 2 = 69.2; P = 0.071). Twenty observational studies were involved in assessing the relationship between AGA and cancer incidence, including 6 cohort studies [The pooled results indicated that any kind of baldness had no influence on the PC-specific mortality in a random-effect model, with an HR = 1.07 (95% CI 0.43–2.64) and moderated heterogeneity  with little evidence of heterogeneity . A summary of the results was presented in Table Four case-control studies , 29, 33 ity Fig. . When thP = 0.038)  Fig. . The effP = 0.029).Based on the Hamilton-Norwood Scale, the 8 study subsets , 33, 35 A total of two studies evaluated AGA at reference age 20 , 34, fivThree studies , 28, 32 As controls, five studies , 34, 35 2 = < 0.1%). In other subgroup analysis, the results were consistent.When four cohort studies , 23, 24 Sensitivity analysis confirmed that no individual study influenced the overall results (data not shown). There was no evidence of publication bias in this meta-analysis indicated by the Begg’s funnel plot and Egger’s tests Fig. .Fig. 3BeThis systematic review and meta-analysis involved a total of 10,935 cases from 21 observational studies. Amoretti et al.  conducteCases of AGA demonstrated a 31% decrease in the incidence of TGCT compared to those with no baldness. The sample size was not sufficient as there were only four studies included in the subset but little heterogeneity was shown. AGA may be regarded as a surrogate of cumulative androgen status which was considered as the result of its components such as exogenous hormones, endogenous hormonal levels, hormonal metabolism, or individual sensitivity to hormones. TGCT is significantly different from PC at the onset age. High risk for TGCT often occurs at ages 20–45 while it occurs at later ages for PC. It has been reported that circulating testosterone, DHT, and E2 declined gradually during male aging . Thus, aFor 11 of the included studies, the association of AGA with PC was not evident in the overall population, which was consistent with the previous study . Some auWhen it came to different categories of baldness, the results showed that the vertex pattern with/without frontal baldness was related to PC risk, but the frontal pattern was not. On the other side, baldness at 2nd stage which amounted to frontal baldness was related to TGCT risk, while the other categories were not. As a result, a dose-response relationship could not be obtained. If these relationships were real, they may be possibly explained by non-linear correlations between the degree of AGA and the circulating androgen status. It was difficult to explain how the results turned to be negative in the PC, when vertex without frontal baldness and vertex with frontal baldness were regarded as independent patterns in the pooled studies. Also, no specific pattern of AGA showed any link to PC risk in the cohort studies. The pathophysiological difference among the patterns of AGA was yet unknown, so a reasonable explanation could not be provided. In addition, the reference age of AGA was different in the included studies; however, Muller et al.  found thIn this meta-analysis, only two of the included studies examined the association of AGA and PC-specific mortality, which showed that inner-study heterogeneity was moderate. Hence, more research work was needed to confirm these findings.Nevertheless, several limitations of this study must be acknowledged. First, despite several subgroup analyses to be performed, significant heterogeneity was generally observed. Given the differences of the studies in race, age, participants’ lifestyle, information collection method, sample size, duration of follow-up and so on, heterogeneity was not avoidable. Second, the number of stratified analyses was so limited that might result in invalid statistical analyses in those groups. Third, although most of the studies used multivariate statistical models to calculate the estimated RRs, the number and content of the adjusted confounders varied in each trial, which might lead to imprecision in the results. But, the most multivariable adjusted-effect estimates were chosen for analysis to minimize the confounding biases. Besides, several sources of bias, such as inherent limitations, unmeasured confounding, and the typical bias of observational studies, could have affected the observed results. Therefore, well-designed and more comprehensive studies are still needed to further evaluate the relationship between AGA and the risk of cancers.The results support the hypothesis that AGA is associated with a reduction of TGCT incidence by altering testicular development and they also suggest that there is a common pathogenic pathway. For PC, positive results are only observed in vertex patterns of AGA and high grade PC, while the viewpoint that AGA can be used as a phenotypic marker for PC risk is poorly supported. In this sense, future studies should be conducted to confirm the conclusions, as well as to evaluate the potential value of this association, which may offer a reference for establishment of predictive models.Additional file 1:Table S1. Review methodology for meta-analysis. (DOC 29 kb)Additional file 2:Figure S1. Male balding patterns base on the Hamilton-Norwood scale. (TIFF 356 kb)Additional file 3:Table S2. NOS scores of cohort studies. Table S3. NOS scores of case control studies. (DOC 86 kb)Additional file 4:Excel S1. Extracted raw data in the meta-analysis. (XLSX 16 kb)Additional file 5:Table S4. Meta-analysis results of association between androgenic alopecia and incidence of cancer. (DOCX 17 kb)


A growing body of evidence shows that childhood wheezing may lead to recurrent or persistent symptoms in adulthood, such that persistent wheezing associated with lung function deficits often have their roots in the first few years of life.We summarized information from several prospective cohort studies following children with or without wheezing into adulthood, to estimate the effect of childhood wheezing on adulthood lung function.Medical literatures were searched in the Medline, PubMed, ScienceDirect, Web of Science and Embase databases up to October 31, 2016. The adulthood lung function was selected as primary outcome, and chronic obstructive pulmonary disease (COPD) prevalence was selected as secondary outcome. The meta-analysis was performed with the Stata Version 14.0. A random-effects model was applied to estimate standardized mean difference (SMD) of lung function, and relative risk (RR) of COPD.Six articles enrolling 1141 and 1005 children with and without wheezing, respectively. Meta-analysis showed that childhood wheezing decreased adulthood lung function as compared with no-wheezing subjects : -0.569~-0.161, P = 0.000). Subgroup analyses indicated that childhood atopic wheezing reduced adulthood FEV1/FVC and FEV1%pred when compared with no-wheezing subjects. In addition, childhood atopic wheezing was significantly associated with COPD prevalence .Our meta-analysis suggests that childhood wheezing may induce ongoing declined lung function that extends into adult life, as well as an increased risk of COPD prevalence when accompanied with atopy. Wheezing is one of the commonest respiratory symptoms in children with a high incidence worldwide. It is manifested by the continuous whistling during breathing, resulting from narrowing or obstruction of the respiratory tract, the two most frequent causes of wheezing in children are clearly asthma and bronchiolitis, the less frequent causes include congenital anatomical abnormalities, aspiration of foreign bodies, and other pulmonary disorders. EpidemiMedical literatures were obtained by searching the following databases: PubMed, Medline, Science Direct, Web of Science and Embase up to 31st October, 2016. The search keywords were: (1) wheeze or wheezing; (2) children or childhood; (3) lung function or pulmonary function. Articles were acquired from PubMed using the following search terms: (wheeze [Title/Abstract] OR wheezing [Title/Abstract]) AND (children [Title/Abstract] OR childhood [Title/Abstract]) AND (lung function [Title/Abstract] OR pulmonary function [Title/Abstract]). Furthermore, additional references were identified from citations in the articles that were reviewed.The related literatures were evaluated by reviewing the titles and abstracts, and further assessed by reviewing the full texts. The studies selected for meta-analysis included those in which data collection began in the childhood, with long-term follow-up into adulthood, and with special focus on serial lung function data. We only included literatures meeting the following criteria: 1) study design: prospective studies must have followed children until 18 years old or above; 2) subjects: studies that included children as the majority of participants were included, for the purposes of the meta-analysis, and age was defined as ranging 0 to 16 years old; 3) wheezing: studies that selected children with wheezing as an observation group and children without respiratory symptoms as a control group; 4) outcome reporting: mean (Any literature was excluded if it belonged to: 1) review, case report, and meeting abstract; 2) repeat publication; 3) non-English article; 4) low quality literature.Two investigators  independently extracted all data, and a third investigator (Xin Peng) settled the differences. The following information was extracted from each article: first author name, year of publication, country, sample size (those subjects that provided lung function data), age of subjects at baseline, duration of follow-up (years), wheeze patterns in each article, assessment method of wheeze, adulthood lung function index , and COPD definition. The Newcastle-Ottawa Scale (NOS) was used for the quality evaluation . A artic2 statistics . A random-effects model was used when significant heterogeneity was observed; otherwise, a fixed effects model was utilized [The meta-analysis was conducted using the Stata Version 14.0. We used standardized mean difference (SMD) to estimate the pooled effect of childhood wheeze on adulthood lung function, and risk ratios (RR) to estimate the association between childhood atopic wheezing and COPD. Precision of the estimates was expressed as the 95% confidence interval (CI), and statistical significance was set at a P value < 0.05 for all analyses. The statistical heterogeneity was assessed using the Q and Iutilized . FurtherWe initially identified 522 references by electronic and manual search. After screening for the titles and abstracts, 434 records were excluded, and 88 full-text articles were selected. The reasons for further exclusion included: review articles (n = 18), adult studies (lacking children) (n = 5), non-prospective studies (n = 12), follow-up not long enough (n = 19), papers provided insufficient data (n = 18), duplicate reports from the same population (n = 3), and low quality literatures(n = 7). Six articles that fulfilled all the criteria were finally included .The six included articles –22 were The scores of included articles ranged from six to eight , which indicated the high quality of included studies and enhanced the reliability of the analysis .2 = 82.1%). Therefore, a random-effects model was used for the pooled analysis. The meta-analysis indicated that adulthood lung function was decreased in childhood wheezing group when compared with control group   .Sensitivity analysis showed that removal of each single study did not alter the overall results of pooled analyses . As show2 = 73.8%)  .There appeared to be funnel plot asymmetry for adulthood lung function , but theWheezing is one of the commonest respiratory symptoms in children with increasing prevalence in recent years. As mentioned, persistent wheezing with lung function deficits often have their roots in the first few years of life, and childhood atopic wheezing raises the risk of COPD in later life. The primary long-term outcome of this meta-analyses is adulthood lung function. 15 prospective cohort studies, which were analyzed, enrolled children with or without wheezing and followed them into various ages of adulthood. Our meta-analysis demonstrated that childhood wheezing decreased adulthood lung function, when compared with no-wheezing subjects. Two mechanisms of adulthood lower lung function were hypothesized, containing the initial damage sustained in early childhood and ongoing pathological progress of pulmonary parenchyma. Pulmonary disease in early life may interfere with lung and airway development . The BAMOur subgroup analysis was based on the pathogenesis of wheezing (atopic or infection-related wheezing) and lung function index . The results demonstrated that childhood atopic wheezing reduced adulthood FEV1/FVC and FEV1%pred when compared with no-wheezing subjects, while no significant differences were identified between the infection-related wheezing patients and controls. We presumed two possible explanations of this result. On the one hand, two published reviews suggested that early airway exposure of atopy was indicated as the most important risk factor to wheezing persistence ,12. The Lastly, the BAMSE cohort  indicateTo the best of our knowledge, this is the first meta-analysis to estimate the effects of childhood wheezing on adulthood lung function. Our meta-analysis was rigorously performed and reported according to the PRISMA checklist. A highly sensitive and comprehensive search of the literature was used in order to identify as many relevant studies as possible and to reduce potential publication bias. We have exclusively collected the studies of prospective design and of high quality. It demonstrated that the childhood wheezing accompany with atopy, as manifested by early airway dysfunction, is associated with progressive lung function decline into adult life and an increased risk of COPD. Nonetheless, our analysis has several unavoidable limitations. The first one is the small number of included articles and the high heterogeneity of lung function, despite our best efforts to retrieve all related data. This is because most literatures failed to meet our stringent inclusion criteria. The high heterogeneity may result from the different degrees of severity of wheezing and huge range of age (from 1 week to 15 years). Unfortunately, the kind of details needed for stratifying the data into further subgroups were not provided in the original studies. Additionally, subjects with COPD in our collected publications were relatively few, also relatively young, ranging from 50 to 65 years old, whereas COPD often develops later in life, and thus, we may have missed potential would-be COPD patients. Clearly, further research, comprising of larger cohort size, more detailed age-related data, and inclusion of older people in longitudinal studies relating childhood wheezing with COPD, need to be undertaken.In general, this meta-analysis suggests that childhood wheezing may induce ongoing declined lung function that extends into adult life, as well as an increased risk of COPD prevalence when accompanied with atopy.S1 Table(XLSX)Click here for additional data file.S1 File(DOC)Click here for additional data file.


ARLTS1) might be associated with an increased risk of several types of familial cancers. However, previous studies have shown that cancer susceptibility is not completely consistent with ARLTS1 polymorphisms, and the precise mechanism remains unknown. Therefore, we conducted a meta-analysis of case-control studies by searching the PubMed, Embase, OVID, Science Direct and Chinese National Knowledge Infrastructure (CNKI) databases. In total, 12 studies met the inclusion criteria and were included in this meta-analysis. Statistical analyses were performed using STATA 11.0 software. Overall, the Cys148Arg T > C variant significantly increased cancer risk . The stratification indicated that the Cys148Arg variant is significantly associated with sporadic cancer  and familial cancer . Trp149Stop, Pro131Leu, Ser99Ser and Leu132Leu were not correlated with cancer susceptibility. Based on these results, we demonstrated that the ARLTS1 Cys148Arg polymorphism is associated with an increased risk of sporadic cancer and familial cancer, and there were no associations between the other four SNPs  and cancer risk.Adenosine diphosphate (ADP)-ribosylation factor-like tumour suppressor gene 1( ARLTS1), which is also known as ADP-ribosylation factor-like protein 11 (ARL11), islocated in chromosomal region 13q14.3 and is a member of the Alternative Reading Frame (ARF) family of the RAS super family of small GTPases, which is involved in apoptotic signalling [ARLTS1 comprises two exons, its second exon encompasses the entire open reading frame andencodes 196 amino acid guanine-nucleotide-binding proteins that are critical components of several different eukaryotic vesicle trafficking pathways [Adenosine diphosphate (ADP)-ribosylation factor-like tumour suppressor gene 1  databases were performed (from January 2005 to November 2016). The following search words and their combinations were used: (“genetic polymorphism” or “polymorphism” or “SNP” or “gene mutation” or “genetic variants”) and  and “ARLTS1 variants  and cancer susceptibility; and (c) the studies had sufficient published data to calculate the odds ratio (OR) with a 95% confidence interval (CI). The primary reasons for the exclusion of studies were as follows: (a) no control population was included, (b) the studies contained overlapping data, and (c) the genotype frequency datawere not available.The studies included in this meta-analysis met the following criteria: (a) the studies used a case-control design; (b) the studies evaluated the association between p-value. The different cancer types were classified as sporadic tumours or familial tumours. The source of the controls was stratified into hospital-based studies and population-based studies.Two investigators independently collected the data from all eligible publications according to the inclusion and exclusion criteria. Whenever disagreements occurred, a third investigator was consulted to resolve the dispute, and a final decision was made by the majority of the votes. The following data were extracted from each study: the first author’s last name, publication year, ethnicity of the study population, study design, cancer type, total number of cases and controls and the Hardy-Weinberg equilibrium (HWE) The quality of the included publications was assessed using a quality assessment scale Table , which wARLTS1 polymorphisms and cancer risk according to a pooled OR with the corresponding 95% CIs. The stratification analysis was performed by type of cancer, control source and study quality. The significance of the pooled ORs was determined by a Z test, and P < 0.05 was considered statistically significant.In our meta-analysis,we investigated the association between χ2 test (significance was set as P < 0.05). The heterogeneity of the studies was examined using the Q-test and the I2 test [P < 0.05 for the Q-test and I2 > 50%), we used a random-effects model (DerSimonian-Laird method) [p-values were two-sided. All analyses were calculated using STATA Version 11.0 software.In our study, the HWE was tested using the  I2 test , 13. If  method) ; otherwi method) . The sen method) . In addi method) . All p-vARLTS1 polymorphisms and cancer susceptibility, were included in this meta-analysis. The characteristics of all studies are summarized in Table p-value.By searching the databases, 31 abstracts were collected according to the search criteria. Of these 31 studies, 11 studies were excluded after reviewing the titles and abstracts. Another 10 studies were excluded after reviewing the full texts. The study selection process is shown in Fig. ARLTS1 Cys148Arg (CC vs. TT: 1.27 (1.15–1.41), p = 0.000 Fig. p = 0.026 Fig. p = 0.045). The meta-analysis results are listed in Table ARLTS1 Cys148Arg CC genotype increased the cancer risk in a population-based (CC vs. TT: 1.38 (1.22–1.55), p = 0.000; CC + TC vs. TT:1.16, p = 0.003; CC vs. TC + TT:1.26(1.07–1.49), p = 0.000), but not in a hospital-based setting. Regarding the study quality, the results were only significant in the high-quality studies (CC vs. TT: 1.35(1.16–1.57), p = 0.001; CC + TC vs. TT: 1.14 (1.00–1.31), p = 0.008). In addition, positive results were observed in both familial cancer (CC vs. TT: 1.26 (1.12–1.43), p = 0.000; CC + TC vs. TT: 1.15 (1.02–1.29), p = 0.021) and sporadic cancer (CC vs. TT: 1.36 (1.18–1.55), p = 0.000; CC + TC vs. TT: 1.13 (1.02–1.25), p = 0.016; CC vs. TC + TT: 1.22 (1.02–1.46), p = 0.030).Among the 12 eligible studies, 11 studies included in this meta-analysis involved SNP C148R with 7152 cases and 6698 controls. Overall, when the 11 studies were pooled into the meta-analysis, the significant main effects on cancer risk were associated with p = 0.700; Pro131Leu CT vs. CC: 0.94 (0.81–1.08), p = 0.374; Ser99Ser AG vs. AA: 1.02 (0.86–1.20) p = 0.847; Leu132Leu GC vs. GG: 0.90 (0.58–1.39) p = 0.845), which are referred to in Table All studies were pooled in the meta-analysis. No significant results were observed for any of these SNPs (Trp149Stop GA vs. AA: 1.11 (0.66–1.86), : 0.90 0.8–1.39 p ARLTS1 Cys148Arg (p = 0.595 for CC vs. TT)  in the gene region. The reduction or absence of ARLTS1 expression has been reported to contribute to DNA mutations with an LOH in breast cancer and a mutation with methylation in thyroid cancer [ARLTS1 polymorphism was not only associated with sporadic cancer but also with familial cancer [ARLTS1 polymorphisms and cancer risk, particularly familial cancer, has recently become a high priority research topic.poptosis . However in lung  and ovar in lung . The expe cancer  and chroe cancer . The potd cancer . This dol cancer . ConsequARLTS1 polymorphisms and cancer risk based on 12 case-control informative studies. Regarding Cys148Arg, we found that the SNP C148R polymorphism was significantly associated with cancer in all subjects. When a subgroup analysis was carried out based on the size of the total sample, we observed different associations with sporadic cancer and familial cancer, which was consistent with the result obtained by Calin et al [ARLTS1 Cys148Arg CC genotype may significantly contribute to decreased apoptosis, thereby promoting the overall cancer incidence. However, the other variants may still have associations with cancer. These variants may have minor effects individually, but they are likely to contribute adequately in combination to lead to the failure of theimmune response. Therefore, gene-gene interactions should be considered in future studies to elucidate the precise role of ARLTS1.To the best of our knowledge, the present study is the first systematic review to investigate the relationship between the Some limitations of this meta-analysis should be addressed. First, this study did not analyse the potential and gene-environment interactions due to a lack of original data, such as data involving environmental risk factors and genotypes. Second, the controls were not uniformly defined. Although most of the controls were selected mainly from healthy populations, some controls had a benign disease. In addition, our results were based on unadjusted estimates without adjusting for other risk factors, such as age, smoking status, drinking status, obesity and environmental factors. Therefore, a more precise analysis should be conducted if individual data were available. Although this study has several limitations, as previously stated, this study is the first to reveal the association between Cys148Arg and cancer risk and exclude the association between the other four SNPs  and cancer risk. Additionally, our study provides powerful evidence for future large-scale population-based cohort and case-control studies.ARLTS1 Cys148Arg polymorphism was associated with sporadic cancer risk and familial cancer risk. However, well-designed studies with larger sample sizes should be performed to confirm this association. Moreover, further studies estimating the effect of gene-gene and gene-environment interactions may eventually provide a better and more comprehensive understanding to clarify the role of ARLTS1 in the pathogenesis of cancer.In summary, the