Inflammation plays a critical role in the development and progression of cancers. The advanced lung cancer inflammation index (ALI) is thought to be able to reflect systemic inflammation better than current biomarkers. However, the prognostic significance of the ALI in various types of cancer remains unclear. Our meta-analysis aimed to comprehensively investigate the relationship between the ALI and oncologic outcomes to help physicians better assess the prognosis of cancer patients.The PubMed, Embase, Cochrane Library, China National Knowledge Infrastructure, and Wanfang databases were searched for relevant studies. Hazard ratios (HRs) with 95% confidence intervals (95% CIs) were calculated and pooled from the included studies. Furthermore, a sensitivity analysis was performed to evaluate the reliability of the articles. Finally, Begg’s test, Egger’s test, and the funnel plot were applied to assess the significance of publication bias.P < 0.001). Moreover, results from subgroup meta-analysis showed that the ALI had a significant prognostic value in non-small cell lung cancer, small cell lung cancer, colorectal cancer, head and neck squamous cell carcinoma, and diffuse large B cell lymphoma .In total, 1736 patients from nine studies were included in our meta-analysis. The median cutoff value for the ALI was 23.2  in the analyzed studies. The meta-analysis showed that there was a statistically significant relationship between a low ALI and worse overall survival (OS) in various types of cancer (HR = 1.70, 95% CI = 1.41–1.99, These results showed that a low ALI was associated with poor OS in various types of cancer, and the ALI could act as an effective prognostic biomarker in cancer patients. Cancer is one of the major causes of death worldwide . In 2018Growing evidence indicates that cancer-related inflammation plays a critical role in the development and progression of various types of cancer –8. At thCachexia in cancer patients is the result of the chronic systemic inflammatory response and often indicates a poor outcome for cancer patients , 18. SarHowever, a pooled study that analyzes the association between the ALI and clinical outcomes of patients with malignant diseases has not been systematically performed. Our meta-analysis aimed to explore the prognostic impact of the ALI in cancer patients, helping physicians predict clinical outcomes more effectively and easily and assisting them in the timely adjustment of therapeutic regimens, which further reduces mortality.This study was performed according to the recommendations of the Preferred Reporting Items for Systematic Reviews and Meta-Analyses statement Studies examining the association between the ALI and prognosis in patients with any type of cancer.(ii)Sufficient data provided to calculate hazard ratios (HRs) and 95% confidence intervals (95% CIs) for the relationship between the ALI and overall survival (OS) in cancer patients.(iii)The cutoff value of the ALI was clear.(iv)If more than one article referred to the same population, only the study that included the most cases or the latest publication was included.The following criteria were used for inclusion in this meta-analysis:(i)Studies based on animal or cell experiments(ii)Meta-analyses, reviews, case reports, or reports based on expert experienceThe following studies were excluded from the meta-analysis:(i)Basic information, including authors’ names, publication year, cancer type, country, study period, characteristics of the study population , survival type, treatments, clinical stage, cutoff value, cutoff selection, and study design(ii)Statistical indicators, including HRs and corresponding 95% CIs for OS, from multivariate or univariate analysis or estimated from Kaplan-Meier survival curves using previously described methods if the HR could not be obtained directly Two authors (HX and HSH) independently extracted the following data from all included studies:The Newcastle-Ottawa Scale (NOS) was used to assess the quality of included studies, and a score ≥ 6 was considered an indicator of a high-quality study, whereas a score < 6 indicated a low-quality study . Two revQ test and I2 statistics were used to analyze heterogeneity between studies; P < 0.05 or I2 > 50% suggested significant heterogeneity among the included studies. If the homogeneity was significant, a random effects model was used. Otherwise, a fixed effects model was used [P values were two-sided, and P < 0.05 was considered statistically significant.Stata software  was used to analyze the data in our study. HRs and 95% CIs were used to evaluate the association between the ALI and OS in cancer patients. A pooled HR > 1 was regarded as an indicator of poor prognosis in groups with a low ALI. The impact of the ALI on survival was considered statistically significant if the corresponding 95% CI for the summary HR did not overlap 1 unit. The Cochran’s was used . SubgrouThe process of study selection is shown in Fig. For quality assessment, the nine studies were evaluated using the NOS, and the scores were all ≥ 6, indicating that the included studies were all high-quality studies . Considering that heterogeneity was not obvious among the studies, a fixed effects model was applied.As shown in Fig. P < 0.001), small cell lung cancer (SCLC) , colorectal cancer , head and neck squamous cell carcinoma , and diffuse large B cell lymphoma . In terms of patient age and sample size, the ALI had a significant prognostic value for cancer patients regardless of the median patient age (≥ 60 years and < 60 years) or sample size (≥ 170 or < 170) . When studies were divided into those performed in Asian, North American, and European countries, the ALI was significantly related to OS only in studies from Asia and North America (P < 0.001 for both). When tumor stage was considered, the results showed that a lower ALI was a risk factor in patients with metastatic or mixed-stage tumors (P < 0.001 for both) but not in patients with non-metastatic disease. When performing subgroup analysis by treatment type, the association was still significant in patients who did not undergo surgery and those who underwent surgery . Furthermore, the ALI was indicated to be an effective prognostic factor when the cutoff for the ALI was > 23.2 and < 23.2 .In addition, subgroup analyses stratified by the median age, sample size, ethnicity, pathological type, clinical stage, treatments, and cutoff for ALI were also performed Table . The resSensitivity analysis was used to detect the robustness of these results, which showed that the pooled results were not altered by any one study, indicating that our conclusions are relatively reliable Fig. .Fig. 3SP = 0.048) and Egger’s test (P = 0.014) indicated that publication bias was present among the studies. An asymmetric funnel plot also proved this conclusion . Furthermore, we observed consistent results in subgroups of various cancer types, including NSCLC, SCLC, colorectal cancer, head and neck squamous cell carcinoma, and diffuse large B cell lymphoma. In short, the ALI could act as a predictive factor for clinical outcomes in cancer patients.A comprehensive search was conducted for published articles exploring the prognostic effect of the ALI on the survival outcomes of cancer patients. A total of 1736 cases from nine studies were included in our meta-analysis. The results of our study indicated that a low ALI was associated with worse prognosis  and a small sample size. Second, the BMI, ALB, and NLR, which are components of the ALI, seem to have better prognostic value in advanced stages of cancer –60. TherOur study inevitably had some limitations. First, all of the studies included in this meta-analysis were retrospective, and the results may have thus been subject to potential bias. Second, confounding factors, such as the levels of tumor markers and history of chemoradiotherapy, might also affect the HR of the ALI in cancer patients; such an effect cannot be explored via subgroup analysis because the studies that were included did not provide sufficient information. Third, the cutoff value of the ALI was not uniform in different studies. Finally, publication bias existed in the studies that were included in our meta-analysis, which may be attributable to failure in publishing studies with negative results or with other variables.In summary, our study revealed that a low ALI was significantly correlated with worse OS in cancer patients. Therefore, the ALI could be a reliable predictor for prognosis in cancer patients, providing consistent results for different cancer types. In the future, more large-scale, prospective, well-designed studies are needed to verify the association of the cutoff values of the ALI and tumor stage with the prognostic features of the ALI for patients with different types of cancer.Additional file 1. PRISMA 2009 checklist used in this meta-analysis.Additional file 2. Newcastle-Ottawa Scale for quality assessment.


I2 = 35.2%, p = 0.117), but a marginal positive association was observed in three cohort studies (relative risk (RR) 1.22, 95% CI: 1.00–1.49, I2 = 0.0%, p = 0.995). Particularly, a negative association was observed in North America . In addition, a positive association was observed in cereal fiber  and a negative association was observed in vegetable fiber . In conclusion, negative associations with endometrial cancer risk were observed for higher total dietary fiber intake and higher vegetable fiber intake in the case–control studies. However, results from the cohort studies suggested positive relationships of higher total fiber intake and higher cereal fiber intake with endometrial cancer risk.Epidemiological studies are inconclusive regarding the association between dietary fiber intake and endometrial cancer risk. Thus, we aimed to conduct a meta-analysis to clarify the association between dietary fiber and endometrial cancer risk. We searched the PubMed and ISI Web databases for relevant studies through March 2018. The association between dietary fiber and endometrial cancer risk was evaluated by conducting a meta-analysis including 3 cohort and 12 case–control studies. A significant negative association was observed between total dietary fiber intake and endometrial cancer risk in 11 case–control studies (odds ratios (OR) 0.76, 95% confidence interval (CI): 0.64–0.89,  Endometrial cancer, ranking the fourth in incidence of all cancers in women, is the most common gynecologic cancer, especially in economically developed countries . Long-laMany studies have suggested that dietary fiber may decrease the risk of multiple cancers, such as colorectal cancer, breast cancer, and pancreatic cancer ,4,5. SevIn 2007, Bandera et al.  reportedThis meta-analysis was planned, conducted, and reported according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) recommendations  and Metahttp://www.wcrf.org), the general search terms of exposure for PubMed included diet [tiab] OR diets [tiab] OR dietetic [tiab] OR dietary [tiab] OR eating [tiab] OR intake [tiab] OR nutrient* [tiab] OR nutrition [tiab]. Specifically for dietary fiber, the key words were consistent with published in Bandera’s meta-analysis [According to WCRF (World Cancer Research Fund) Specification Manual  case–control studies, prospective studies, or clinical trials; (2) that studied dietary fiber and other fiber subtype exposure; (3) that focused on endometrial cancer as an outcome; and (4) reported the odds ratio (OR) or relative risk between the intake of dietary fiber and the risk of endometrial cancer. Investigators identified studies based on titles, abstracts, and full texts independently. Thirty eligible articles were retrieved after titles and abstracts were reviewed. As per the exclusion criteria, (1) review articles; (2) studies based on cell lines and animals; (3) studies not related to dietary fiber and endometrial cancer risk; and (4) studies which did not assess ORs or RRs were removed. After going through the full text, 1 meta-analysis and 14 studies which did not assess dietary fiber as an exposure variable were excluded in the 30 eligible articles. Finally, 15 articles were included in this meta-analysis . Data was extracted from each study, including last name of the first author, year of publication, country, age, sample size, number of cases, method of dietary assessment , exposure (dietary fiber and other fiber subtypes), quantity of intake, hysterectomy, effect size (odds ratios (ORs) for case control study, relative risks (RRs) for cohort study), 95% confidence intervals, and adjusted variables. We presented evidence from all the selected studies, and then repeated certain analyses, excluding studies that did not meet certain quality criteria. The criteria were (1) case–control studies or cohort studies; (2) sample size of at least 200 cases for more optimal statistical power; (3) exclusion of hysterectomies from the control group; and (4) adjustment for important confounders, such as total energy and body mass index.According to the WCRF criteria , there were enough studies to conduct meta-analysis for total dietary fiber and cereal fiber. Although vegetable fiber, fruit fiber, and insoluble fiber did not meet the WCRF criteria, we also performed the meta-analysis for studies with vegetable fiber, and provide a table reporting the data of studies with fruit fiber and studies with insoluble fiber.I2 statistics were used to examine statistical heterogeneity among studies. As long as the p value for heterogeneity was less than 0.05 or the I2 statistic was more than 25%, results of the studies were considered as significantly heterogeneous and a DerSimonian and Laird random effects model was used to pool the results [P value was less than 0.05, the trim and fill method was used to adjust the publication bias [The squared inverse variance for the logarithm RR/OR was considered as the appropriate weight for each study. The Q test and  results ,16. Publion bias .We used DerSimonian and Laird random effects models to calculate summary ORs, RRs, and 95% confidence intervals (CIs) for the highest versus the lowest levels of total dietary fiber, cereal fiber, and vegetable fiber. Our meta-analysis excluded the data from Shu et al.  because All statistical analyses were performed with STATA version 12.0 . All statistical tests were 2-sided.There were 3 cohort studies, 10 population-based case–control studies, and 2 hospital-based case–control studies identified for the current meta-analysis . Overallp = 0.315) and the Begg’s test (p = 0.324) suggested no significant publication bias . Based on the three cohort studies, a marginal positive association was observed between dietary fiber intake and endometrial cancer risk . Based on the three cohort studies, a significantly positive association was observed . However, the summary ORs of three case–controls indicated a negative association of vegetable fiber intake with endometrial cancer risk  were included in the analysis of fruit fiber and risk of endometrial cancer. In total 4 studies (1 cohorts and 3 case controls) were included in the analysis of insoluble fiber and risk of endometrial cancer. Only Goodman et al.  found a Our meta-analysis supports an inverse association between total dietary fiber intake and risk of endometrial cancer in case-control studies, particularly in studies conducted in North America. The result of cohort studies for total fiber was inclined to a positive direction. The result of cohort studies for cereal fiber was significantly positive. Only three case–control studies support an inverse association with vegetable fiber. No significant association was observed for intake of fruit fiber and insoluble fiber. Our results for total dietary fiber derived from case–control studies is consistent with previous meta-analysis . HoweverAdults who consumed large amounts of dietary fiber, compared with those who have moderate fiber intake, were at lower risk for obesity, hypertension, diabetes, cardiovascular diseases, and gastrointestinal diseases . RecentlIt is a challenge to expound the biochemical mechanisms for total dietary fiber and endometrial cancer risk, for the complex components of total dietary. These analyses were based upon questionnaires in traditional epidemiologic studies, and dietary fiber was defined in different standards in questionnaires. However, substantial research has been conducted to evaluate the effect of dietary fiber and body weight, most of which showed an inverse relationship between dietary fiber intake and change in body weight . WCRF anOn the other hand, bile acids (metabolites of cholesterol in liver) promote fat absorption in the intestinal, tract which is then reabsorbed to the liver by enterohepatic circulation. Dietary fiber mechanically binding bile acids results in less reabsorption of bile acids, and thus, reduces cholesterol in plasma, which is a precursor of the endogenous synthesis of estrogens. Meanwhile, absorption of fats is also restrained. Lower fat intake and lower adiposity are associated with lower endometrial cancer risk. A high-fiber diet is a healthy dietary pattern associated with lower endometrial cancer risk.Aune et al. supported a negative association between cereal fiber and colorectal cancer risk . InteresOur meta-analysis suggested an inverse association between vegetable fiber and endometrial cancer risk, but no significant result was found for fruit fiber and insoluble fiber. Although three case–control studies support the null results, vegetable fiber and fruit fiber appeared to be protective. However here we should consider the same questions on how to accurately evaluate the biochemical constituents in these subtypes of fiber. Biochemically, dietary fiber is composed of cellulose, hemicellulose, lignin, pectin, and beta-glucans. These components are resistant to digestion in small intestine and require bacterial fermentation in large intestine ,8,35. LiSoluble fiber contains pectin, gums, inulin-type fructans, and some hemicelluloses, which have a relatively lower glycemic load. These characteristics are associated with favorable effects on glucose and insulin metabolisms and insulin resistance. Soluble fiber has also been shown to result in lower blood pressure levels and diabetes risk. Both hypertension and diabetes are risk factors for endometrial cancer . UnfortuCellulose, pectin, and beta-glucans are polysaccharides that bypasses enzymatic digestion of small intestines, but are easily degraded by microflora in colon. The gut microbiome can increase circulating hormones via deconjugation and influence the composition (or types) of hormones in circulation via hydroxylation/dehydroxylation and methylation/demethylation . The comIn addition, bacterial fermentation of fiber leads to production of short-chain fatty acids, which may have protective effects against cancer. Approximately 95% of short-chain fatty acids in the gut are absorbed and metabolized by the host for a wide range of physiological functions. Microbial-generated acetate has been shown to bind a G-protein-coupled receptor, GPR43, expressed in immune cells . This suOur meta-analysis has several strengths. This is the first time that the positive association for cereal fiber intake on endometrial cancer risk was observed by meta-analysis. Besides, we were firstly concerned with geographic effects, and observed the significantly negative association for dietary fiber intake in endometrial cancer risk in North America. It is suggested that the effects of dietary fiber intake could be related with dietary patterns and lifestyles. There are also limitations for this meta-analysis. First, there were not enough studies to explore the association between other fiber subtypes  and endometrial cancer risk. Second, the results are subject to the influence of measurement error linked to the nature of food frequency questionnaire. Precise measurement of dietary fiber biochemical constituents is a challenge to understand the relationship with endometrial cancer risk. With respect to dietary fiber as a type of substrate of bacterial fermentation, it is necessary to precisely measure microflora and its metabolite to explore the relationship of dietary fiber, gut flora, and endometrial cancer risk. Finally, none of these studies analyzed different types of endometrial cancer. Type I endometrial cancer is estrogen-dependent and associated with endometrial hyperplasia, whereas type II endometrial cancer is estrogen-independent and associated with endometrial atrophy .In conclusion, our meta-analysis observed a negative association between dietary fiber intake and endometrial cancer risk in 12 case–control studies and a significantly negative association in North America. Vegetable fiber intake tended to be negatively associated with endometrial cancer risk. Besides, the three cohort studies suggested positive associations of higher total fiber intake and higher cereal fiber intake with endometrial cancer risk. The mechanism of dietary fiber in endometrial cancer prevention warrants further research. Future prospective studies to evaluate the effect of dietary fiber should consider gut flora in relation to the risk of type I and type II tumors in endometrial cancer .


ICAM-1) gene E469K polymorphism is one of the most commonly studied polymorphisms in this gene because of its association with CHD risks, but results were conflicting. The PubMed, Embase, and China National Knowledge Infrastructure databases were searched for case–control studies published up to November 2018. Pooled odds ratios (ORs) and 95% confidence intervals (95% CIs) were calculated to assess the association. Eleven eligible studies, comprising 3435 cases and 3199 controls, were included in the meta-analysis. The pooled result showed that the ICAM-1 gene E469K polymorphism was significantly associated with an increased risk of CHD . Subgroup analysis supported the results in the Chinese populations and in the Caucasian populations. This meta-analysis suggests that the ICAM-1 gene K469E polymorphism is associated with CHD risk and the K allele is a more significant risk factor for developing CHD amongst Chinese and Caucasians populations.Coronary heart disease (CHD) is a complex polygenic disease in which gene-environment interactions play a critical role in disease onset and progression. The Intercellular adhesion molecule 1 ( Coronary artery disease (CAD) continues to be a leading cause of morbidity and mortality amongst adults globally and represents a public health challenge in both industrialized and developing countries . GeneticIntercellular adhesion molecule 1  – a member of the large immunoglobulin superfamily – is widely expressed at a basal level and can be up-regulated by proinflammatory cytokines . It is eICAM-1 gene, resulting in the substitution of lysine to glutamate (K469E), might have possible functional value in the etiology of atherosclerosis [A common genetic polymorphism rs5498 at position 1548 in codon 469 in exon 6 of the A relatively large number of studies have been published on the association of ICAM-1 E469K polymorphisms and CHD risk, but the findings have been inconsistent. It is possible that a single study may have been underpowered for detection of a small effect of the polymorphisms on disease risk, especially when the sample size was relatively small. Different types of study populations and study design may also contribute to the disparate findings. To clarify the effect of the ICAM-1 E469K polymorphism on the risk for CHD, we performed a meta-analysis of all eligible case–control studies that have been published.Since our study was a meta-analysis based on published articles, we did not draft a statement of patient consent or seek the approval of internal review boards. The electronic databases of PubMed, Embase, and China National Knowledge Infrastructure were searched for studies to include in the present meta-analysis, by using the terms ‘ICAM-1’, ‘intercellular adhesion molecule-1’, ‘polymorphism’, ‘CHD’ and ‘ischemic heart disease’, ‘ACS’, and ‘myocardial infarction’. An upper date limit of 25 November 2018 was applied; we used no lower date limit. The search was done without restriction on language but was limited to studies that had been conducted on human subjects. We also reviewed the Cochrane Library for relevant articles. The reference lists of reviews and retrieved articles were hand searched simultaneously. Only published studies with full text articles were included. When more than one of the same patient population was included in several publications, only the most recent or complete study was included in this meta-analysis.The included studies met the following criteria: (i) evaluated ICAM-1 K469E polymorphism and CHD risk; (ii) case–control studies; (iii) supplied the number of individual genotypes in cases and controls; and (iv) indicated that the distribution of genotypes amongst controls were in Hardy–Weinberg equilibrium.Information was carefully extracted from all eligible publications independently by two authors according to the inclusion criteria listed above. Disagreement was resolved by discussion between the two authors. The following data were collected from each study: first author’s surname, year of publication, ethnicity (country of origin), total numbers of cases and controls, and numbers of cases and controls with K, E, E/E, K/E, and K/K genotypes. We did not contact the author of the primary study to request the information. Different ethnicity descent was categorized as Asian and Caucasian. We did not define any minimum number of patients as required to include a study in our meta-analysis.P value greater than 0.10 for the Q-test indicated a lack of heterogeneity amongst studies so that the pooled OR estimate of each study was calculated by the fixed-effects model (the Mantel–Haenszel method) [t-test, as suggested by Egger  [Odds ratios (OR) with 95% confidence interval (CI) were used to assess the strength of association between the ICAM-1 K469E polymorphism and CHD risk. The OR of CHD associated with ICAM-1 K469E genotype, the K allele carriers (K/E + K/K) versus E/E genotype, allele K versus allele E were calculated, respectively. Subgroup analyses were carried out with respect to ethnicity. Heterogeneity assumption was checked by the Chi-square-based Q-test [ method) . Otherwi method) . One-wayA total of 14 publications , involviAcute coronary syndrome (ACS), prior MI, stable angina pectoris cases were included. Appropriate diagnosis criteria and proper genotyping methods were used in most of the studies. There were ten groups of Asians and four groups of Caucasians. Hardy–Weinberg equilibrium had been tested for all polymorphisms in the control subjects and all were found to be in Hardy–Weinberg equilibrium.P = 0.000 for heterogeneity) (P = 0.000 for heterogeneity), when compared with the homozygous wild-type genotype (E/E) (geneity) , for thepe (E/E) .P = 0.005 for heterogeneity) and allele K versus allele E  and amongst Caucasians for both the K allele carriers versus E/E  and allele K versus allele E .In the stratified analysis by ethnicity, significant association was identified amongst Asians for both the K allele carriers versus E/E .Begg’s funnel plot and Egger’s test were performed to access the publication bias of the literature. The shapes of the funnel plots did not reveal any evidence of obvious asymmetry . AccordiICAM-1 is a ligand for lymphocyte function-associated antigen 1 and integrinβ-2, making it an important player in various inflammatory/immune conditions, including atherosclerosis. Evidence implicating ICAM-1 in the pathogenesis of atherosclerosis and CHD come from both human and animal studies. Animal studies have demonstrated that transgenic mice deficient in ICAM-1 expression have reduced fatty streak formation compared with wild-type mice . In 2002Our meta-analysis summarized for the first time all the available data on the association between ICAM-1 K469E polymorphisms and CHD risk, including a total of 14 studies involving 3435 CHD cases and 3199 controls. Our results indicated that the ICAM-1 K469E polymorphisms were significantly associated with an increased risk of CHD; the K allele in the ICAM-1 K469E polymorphisms was determined to be a more significant risk factor for developing CHD in all populations. In the stratified analysis by ethnicity, significant association was identified amongst Chinese and Caucasians. However, the combined OR (1.76) for the Caucasians populations was larger than the combined OR (1.58) for the Chinese, suggesting that the K allele in the ICAM-1 K469E polymorphisms could play an important role in the pathogenesis of CHD in Caucasians. Population stratification is an area of concern and can lead to spurious evidence supporting the association between a marker and a disease, in effect suggesting a possible role of ethnic differences in genetic backgrounds and the environment they lived in. In addition, it also likely that the observed ethnic differences may be due to chance because studies with small sample size may have insufficient statistical power to detect a slight effect or may have generated a fluctuated risk estimate.Meta-analytic methods are powerful tools for studying cumulative data from individual studies with small sample sizes and low statistical power. Pooling the effects from individual studies by a meta-analysis may increase the statistical power and can help detect modest risk differences amongst study groups. The large dataset of this pooled analysis enabled us to investigate the association between ICAM-1 E469K polymorphisms and CHD risk that could not be addressed adequately in previous case–control studies.There are several limitations inherent to meta-analysis that should be considered when interpreting these results. First, heterogeneity is a potential problem when interpreting any results obtained by meta-analyses. We minimized this likelihood by performing a careful search for published studies, using explicit criteria for study inclusion, precise data extraction, and strict data analysis as best as we can. However, some pooled ORs were obtained from heterogeneous studies. Second, only published studies were included in this meta-analysis. The presence of publication bias indicates that non-significant or negative findings may be unpublished. At last, the studies included in the present meta-analysis are case–control studies, not randomized population-based surveys, and may be biased by problems of stratification. It is possible that controls were not recruited from exactly the same genetic population as the CHD patients.In conclusion, the present meta-analysis suggests that the K469E polymorphism of ICAM-1 is significantly associated with an increased risk of CHD, and the K allele appeared to be a more significant risk factor for developing CHD amongst Chinese and Caucasians populations.


In contrast to younger athletes, there is comparatively less literature examining cardiac structure and function in older athletes. However, a progressive accumulation of studies during the past four decades offers a body of literature worthy of systematic scrutiny.We conducted a systematic review, meta-analysis and meta-regression of controlled echocardiography studies comparing left ventricular (LV) structure and function in aerobically trained older athletes (> 45 years) with age-matched untrained controls, in addition to investigating the influence of chronological age.Q and I2 statistic.Electronic databases were searched from inception to January 2018 before conducting a random-effects meta-analysis to calculate pooled differences in means, effect size and 95% confidence intervals (CIs). Study heterogeneity was reported using Cochran’s 2, 95% CI 19.84–36.49) and stroke volume  . Athletes had superior global diastolic function , lower A  and a′ , and more rapid e′ . Meta-regression for chronological age identified that athlete–control differences, in the main, are maintained during advancing age.Overall, 32 studies  were included. Athletes had greater LV end-diastolic diameter , interventricular septal thickness , posterior wall thickness , LV mass , LV mass index  contains supplementary material, which is available to authorized users. It is well established that a lifestyle consisting of regular physical exertion is associated with reduced cardiovascular risk and all-cause mortality across the age spectrum , 2. DespA major limitation of this area, however, is that randomised controlled trials of life-long exercise are highly unlikely, and therefore researchers must rely on evidence from animal models and controlled observational studies to further our understanding of the effects of age on cardiac structure and function , 7–9. InIn light of this, we undertook a systematic review and meta-analysis of studies using echocardiography to (1) compare LV morphology and systolic and diastolic function in older athletes versus age-matched sedentary but otherwise healthy controls and (2) employ meta-regression to explore the influence of chronological age on cardiac structure and function in the ageing athlete.The systematic search processes, evaluation, analysis, and reporting were conducted and presented in accordance with the Preferred Reporting Items for Systematic review and Meta-Analyses (PRISMA) guidelines for reporting systematic reviews and meta-analyses .An electronic database search was designed by two authors (AB and NS), who conducted an independent literature search of PubMed (title and abstract), MEDLINE (title) and ScienceDirect  for published, English-language journal articles from inception to 11 January 2018 .Studies were eligible for quantitative analysis when meeting the following criteria: (1) study participants were male; (2) mean participant age ≥ 45 years; (3) aged-matched control group; (4) athlete groups were aerobic/endurance trained; (5) control groups were untrained; (6) participants were reported as free from cardiovascular diseases; (7) the study was observational in design and data were recorded at a single time point (including recruitment data from intervention studies); (8) studies used echocardiography; and (9) studies assessed cardiac strain using two-dimensional speckle tracking echocardiography (STE).We elected to include only endurance athletes participating in predominantly dynamic aerobic activity . Where study information was unclear, corresponding authors were contacted by email.® Excel 2016, Microsoft Corporation, Redmond, WA, USA). Twenty-one measures of LV, RV and LA structure and function were recorded. Cardiac structure measures included (1) interventricular septum (IVS) thickness, (2) LV posterior wall thickness (PWT), (3) left ventricular end-diastolic diameter (LVEDD), (4) relative wall thickness (RWT), (5) left ventricular mass (LVM), (6) LVM index (LVMi), (7) left ventricular end-diastolic volume (LVEDV), (8) right ventricular end-diastolic diameter (RVEDD), and (9) left atrial diameter (LAD). LV systolic function measures by conventional echocardiography were (10) ejection fraction (EF), (11) fractional shortening (FS), (12) stroke volume (SV), (13) systolic annular tissue velocity (s′), and by two-dimensional STE, (14) global longitudinal strain (GLS). Diastolic function measures were (15) early mitral inflow velocity (E), (16) late mitral inflow velocity (A), (17) ratio of early to late mitral inflow velocity (E/A), (18) early diastolic annular tissue velocity (e′), (19) late diastolic annular tissue velocity (a′), (20) ratio of early to late diastolic annular tissue velocity (e′/a′), and (21) ratio of early mitral inflow velocity to early diastolic annular tissue velocity (E/e′).Following an initial literature search (AB), study data were independently extracted by AB and AC and cross-checked by an arbitrator (NS). Extracted data were entered into a spreadsheet  were combined to obtain a global value. We included studies employing either pulsed-wave Doppler or colour Doppler techniques to obtain tissue velocities. Data for RWT included both descriptions of RWT and thickness/radius ratio. We elected to only include studies that scaled LVM to body surface area (BSA) in line with current echocardiographic recommendations . GLS waspressure .N, where N is the number of participants. Likewise, when median and range were reported, a manual conversion was applied to convert data into mean and SD, in accordance with the study sample size  met the inclusion criteria for quantitative analyses –59. ThisSummarised participant anthropometrics, systolic and diastolic blood pressure and heart rate are presented in Table IVS, PWT, LVEDD, LVEDV, LVM and LVMi were greater in athletes compared with controls, while RWT, RVEDD and LAD did not differ. Significant heterogeneity was observed for all parameters, with inconsistency considered moderate for LVEDD, yet high for IVS, PWT, RWT, LVM, LVMi, LVEDV RVEDD and LAD.s′ and GLS were not different between athletes and controls, whereas SV was greater in athletes. Between-study heterogeneity was significant in all cases and the inconsistency considered moderate in FS and GLS, yet high for EF, s′ and SV.EF, FS, E between athletes and controls, albeit A was significantly lower in athletes. Accordingly, E/A was greater in athletes compared with controls. Tissue velocity analyses showed higher e′ in athletes, whereas athletes showed significantly reduced a′ but greater e′/a′ compared with controls. E/e′ was not different between athletes and controls. Between-study heterogeneity was significant for all parameters of diastolic parameters besides E/e′. Inconsistency was low for E/A and E/e′, moderate for E, A and a′, and high for e′ and e′/a′.Pooled analysis of studies measuring mitral inflow velocity revealed no difference in A .−1 greater cardiac output at rest) than controls, and this difference is sustained with chronological ageing.This first systematic pooling of controlled echocardiographic evidence of cardiac structure and function in aerobically (endurance) trained older athletes compared with age-matched untrained controls reveals that (1) endurance-trained older athletes have superior diastolic function compared with untrained counterparts, which is sustained regardless of advancing age; (2) mean LVM of older endurance athletes is greater (mean difference 72 g) than that of controls, as evidenced by greater wall thickness (mean ~ 1.2 mm) and chamber diameter (mean difference 3.7 mm); (3) despite modest differences in global systolic function between groups, older athletes have a larger SV  compares well with that reported for younger individuals (mean difference 0.2) by Utomi et al. [A, without concomitant changes in E, accords with a large investigation of young, Olympic calibre athletes [A between athletes and controls becomes more exaggerated with advancing age, which suggests an incremental reliance on atrial contraction to support LV filling in untrained controls.Older athletes enjoy more favourable global diastolic function which is driven by a reduced reliance on late diastolic filling and evident in both haemodynamic and tissue assessment. Greater ontrasts  and concontrasts  with prei et al. , suggestathletes , highlige′ in older trained adults agrees with findings from Utomi et al. [a′, and when considered with the observed difference in early diastole, a greater e′/a′ was observed in athletes than in controls. These observations support the precept that continuation of a high volume of aerobic exercise into advanced age has pleiotropic effects on cardiac function.Greater i et al.  in youngOlder athletes presented with larger LV wall thicknesses, absolute and relative LVMs, chamber diameter and LVEDV, which is comparable with findings from large-scale meta-analyses in younger athletes , 13. ThiLAD was similar between athletes and controls, which contrasts with a prior meta-analysis showing larger LAD and volumes in younger athletes than in controls . HoweverSimilarly, RVEDD did not differ between older athletes and controls, which agrees with previous work in younger endurance athletes. Yet, two of the three studies in this meta-analysis observed larger RVEDD in athletes than in controls, highlighting a biventricular enlargement , 35 but Conventional measures of LV systolic function, EF and FS were comparable between athletes and controls, as observed in younger cohorts , 13, whiSpeckle tracking-derived GLS is a more sensitive marker of LV systolic (dys)function than conventionally measured EF . HoweverWithin the present systematic search there was a single study of female athletes . Due to Reporting of athlete training was inconsistent between studies, which prevented additional meta-regression analyses of training years, volume and intensity to further elucidate the between-study heterogeneity. Therefore, more robust documentation with quantitative means ± SDs is required before the influence of training regimes can be documented within a narrower age range, since cardiac structure and function have been reported to be exercise dose dependent . Many st2.7 or fat-free mass to index LVM, which prevented separate meta-analysis and regression using these parameters for indexing other than BSA. It is noteworthy that two studies [This meta-analysis has notable limitations. Firstly, we were confined to analysing male participants. Further, we did not include other measures of body size-adjusted LV morphology . This was due to few data and inconsistent scaling within the literature. Of the included studies for LVMi, five used either height studies , 51 did  studies . This wa studies , 51 and  studies , whereas studies  observedPooled information from controlled echocardiography studies demonstrates that older athletes have superior global diastolic function compared with controls because of a reduced reliance on atrial contraction to LV filling in mitral inflow velocity and both increased early and reduced late diastolic tissue velocities. Furthermore, older athletes have notable differences in cardiac structure  with greater relative and absolute LVM. Despite unremarkable differences in LV systolic function, SV is markedly greater in older athletes. Notably, the present data also identify that masters athletes maintain these functional effects during chronological ageing.Supplementary material 1 (DOCX 13 kb)Supplementary material 2 (DOCX 418 kb)Below is the link to the electronic supplementary material.


Since the late 1990s, hand, foot and mouth disease (HFMD) has become a common health problem that mostly affects children and infants in Southeast and East Asia. Global climate change is considered to be one of the major risk factors for HFMD. This study aimed to assess the correlation between meteorological factors and HFMD in the Asia-Pacific region. PubMed, Web of Science, Embase, China National Knowledge Infrastructure, Wanfang Data and Weipu Database were searched to identify relevant articles published before May 2018. Data were collected and analysed using R software. We searched 2397 articles and identified 51 eligible papers in this study. The present study included eight meteorological factors; mean temperature, mean highest temperature, mean lowest temperature, rainfall, relative humidity and hours of sunshine were positively correlated with HFMD, with correlation coefficients (CORs) of 0.52  0.42–0.60), 0.43 (95% CI 0.23–0.59), 0.43 (95% CI 0.23–0.60), 0.27 (95% CI 0.19–0.35), 0.19 (95% CI 0.02–0.35) and 0.19 (95% CI 0.11–0.27), respectively. There were sufficient data to support a negative correlation between mean pressure and HFMD . There was no notable correlation with wind speed . Our findings suggest that meteorological factors affect the incidence of HFMD to a certain extent. Hand, foot and mouth disease (HFMD) is a common infectious disease caused by a group of enteroviruses, including coxsackievirus A16 (CA16) and enterovirus 71 (EV71) , includiet al. [et al. [et al. [et al. [et al. also reported no significant correlation between mean pressure and the incidence of HFMD. These inconsistencies also existed with respect to other meteorological factors.Although China pioneered the development of an innovative EV71 HFMD vaccine, this vaccine was only effective to prevent HFMD caused by EV71 infection, and could not prevent HFMD caused by infection with other enteroviruses (including CA16) . Therefoet al.  reported [et al.  reported [et al.  found th [et al.  reportedIt was clear from previous studies that the correlation between meteorological factors and HFMD were not consistent. Therefore, the current study was conducted to perform a meta-analysis of published articles, with the aim of identifying meteorological risk factors for HFMD and the role of these risk factors in the pathogenesis of HFMD. The aim is to provide baseline information and a scientific basis for prevention and control measures of HFMD.A literature search was performed using PubMed, Web of Science, Embase, China National Knowledge Infrastructure, Wanfang Data and Weipu Database. We used the following key words: hand foot and mouth disease (HFMD), meteorological, climate change, temperature, precipitation, air pressure, humidity, wind speed and sunshine. Articles published before May 2018 were included in the current study. References from the retrieved documents were also checked to include any additional relevant articles.The inclusion criteria were as follows: (1) article language was Chinese or English; (2) study with a reported sample size; and (3) study provided clear correlation coefficient (COR), risk ratio (RR), odds ratio (OR) and incidence rate ratio (IRR) between meteorological factors and HFMD incidence.Exclusion criteria were as follows: (1) review article; (2) repeatedly published papers; (3) articles with irrelevant data or lack of required information.The first author, year of publication, title, location, study period, time sample size, meteorological factors and COR were recorded on a form. To evaluate the impact of study quality on the results of the study, we designed an assessment programme according to standard guidelines –16. The Q and I2 statistics were used to estimate heterogeneity among studies. According to the Q-statistic, if P value of <0.10 indicates heterogeneity in the risk factors among studies, then in such cases the random-effect model was used for the meta-analysis. Otherwise, the fixed-effect model was used. When we extracted data from the included literature, we found that most of the articles used Spearman correlation to analyse the relationship between meteorological factors and HFMD, for instance, Chen et al. [et al. [R software  was used for meta-analysis. First, n et al. , and the [et al. . If the  [et al. –21, OR [ [et al. ,23 and I [et al. ,21,23–25 [et al. . Then me [et al. . The steπ is the mathematical constant (approximately 3.14).We converted from the log OR to the standardised mean difference (SMD) usingr value) usingA is the parameter related to the sample size (n1 and n2) of the two sets of data in the correlation analysis, where n1 ≠ n2We then could convert from the standardised mean difference (SMD) to the correlation (A) depends on the ratio of n1 to n2, rather than the absolute values of these numbers. Therefore, if n1 and n2 are not sure, using n1 = n2, this will yield A = 4.The parameter  of each study by Fisher's Z transform. We calculated Fisher's Z value and its standard error SEz for analysis, which yield the summary effect (summary Z) and 95% confidence interval (CI). Then the summary Z value was transformed into the summary r value [Summary statistics were then calculated, as most meta-analyses do not directly use CORs when combining CORs, because the variance of each COR is too dependent on the correlation –29. The r value) to Fisher's Z value is given byThe transformation from sample COR  usingThen we convert each of summary r, and 95% CI. Based on the heterogeneity test results, we then judged the combination of effect values (summary r) using a random-effect model or a fixed-effect model. The hypothesis test was used to judge whether the correlation was statistically significant. The data were calculated and transformed using R software. Finally, forest plots were used to indicate the effect size. Publication bias was assessed with funnel plots and Egger's test.To calculated the summary A total of 2437 articles were retrieved and 51 articles were included in the present study . DetaileAll studies shown in I2 = 97%, P < 0. 001) suggested that there was heterogeneity, using the random-effects model to pool the effect values. The combined effect showed that the correlation between mean temperature and HFMD was statistically significant. CORs  were 0.52 (95% CI 0.42–0. 60), −0.51(95% CI −0.63 to −0.36) and 0.10 (95% CI −0.03 to 0.23) (Heterogeneity test: mean temperature (to 0.23) , 5, 9, t95% CI 0.2–0. 60, to 0.23) , respectThe forest plots of other meteorological factors can be found in The study also included subgroup analysis. In I2 = 19.7%, P = 0.003). The studies of tropical climate found no statistically significant correlation between the incidence of HFMD and any of the meteorological factors. The results of subgroup analysis based on exposed time resolution (climate group) were consistent with those without stratification, as shown in In subgroup analysis by regional climate, the studies of subtropical climate showed a slightly lower COR than the non-stratified group in relation to all the meteorological factors except average relative humidity, and the association of hours of sunshine with HFMD was not statistically significant and showed less heterogeneity , that for mean maximum temperature was 0.43 (95% CI 0.23–0.59). CORs for mean minimum temperature, rainfall, mean relative humidity and sunshine were 0.43 (95% CI 0.23–0.60), 0.27 (95% CI 0.19–0.35), 0.19 (95% CI 0.02–0.35) and 0.19 (95% CI 0.11–0.27), respectively. These meteorological factors showed positive correlations with HFMD. However, mean air pressure was negatively correlated with HFMD, while mean wind speed showed no significant association . The mechanism of meteorological factors on HFMD remains unclear. Possible reasons to explain these findings may be that, first, elevated temperatures contribute to the survival, reproduction and transmission of enterovirus in the outside environment . There iet al. [r = −0.36, P = 0.005), a finding that might be accounted for by a gap between the star time of the study and the establishment and improvement of the HFMD surveillance system [et al. [The results of this study were consistent with the majority of research findings. However, a study conducted by Wei et al.  reportede system . In addi [et al.  were notIn subgroup analysis, no statistical significance was reported between mean relative humidity and subgroup of days. The same pattern was reported for hours of sunshine. It is possible that different time scales have an impact on the results of the study; while on the other hand, the effect was non-significant among the larger sample size studies. On the time scale, there was a lag effect in the subgroup of weeks, and there were seasonal problems in the monthly subgroup, all of which should be considered. It was concluded that time scales might be the factors affecting heterogeneity. As shown in The strength of our study is based on its design as a meta-analysis; this was the first meta-analysis to examine correlations between eight meteorological factors and the incidence of HFMD. We observed that some meteorological factors such as temperature, air pressure, duration of sunshine, humidity and rainfall might be risk factors for HFMD and confirmed that these meteorological factors might affect the incidence of HFMD to a certain extent, more reliable and power than the conclusions obtained from single studies. As shown in The study has some limitations. First, we found the significant heterogeneity between the studies included in this meta-analysis. All of the included studies were conducted in Southeast and East Asia, especially in China. So differences in regional and analytical methods might result in high heterogeneity in estimates from the literature. Most of the included studies that were conducted in subgroup analysis only analysed time resolution and regional climate. Additional possibilities such as national income should be considered to better understand sources of heterogeneity. The second limitation is that we only considered the eight meteorological factors affecting HFMD, ignoring other factors. In the present study, there were few studies of some of the meteorological factors involved, such as mean maximum temperature, so the results might be biased. Confirmation and clarification of these findings will require larger sample size and wider research region, such as the studies reviewed were from other Asian countries or throughout the world. Lastly, the studies we examined in this meta-analysis were studies of association. The researches we included were cross-sectional designs that limited the causal inference. The true relationship between meteorological factors and HFMD could be more complex, and there are potential difficulties in accurately measuring complex associations. Therefore, the association between meteorological factors and HFMD might be one of many influencing factors.In summary, among the eight meteorological factors examined, the average temperature, average maximum temperature, mean minimum temperature, mean air pressure, rainfall, mean relative humidity and sunshine were related to HFMD, indicating that these factors play important roles in the incidence of HFMD. However, only eight major meteorological factors were analysed in this study. HFMD is a multifactorial disease that may be affected by additional meteorological factors . Further analyses should examine various comprehensive indicators.In conclusion, the results of this meta-analysis provide epidemiological evidence that meteorological factors (such as temperature and air pressure) may increase the incidence of HFMD in the Asia-Pacific regions. Further research should be performed to explain clearly the correlation between meteorological factors and HFMD in other areas of the world, outside the Asia-Pacific region. At the same time, monitoring these meteorological factors would play a warning role in the occurrence and prevalence of HFMD and could provide information useful in the development of prevention and control measures for HFMD, particularly in subtropical and temperate climates.


As there is no consensus on the optimal surgery strategy for multiple primary lung cancer (MPLC), we conducted this study to address this issue by comparing the prognosis of MPLC patients underwent different surgical strategies including sublobar resection and the standard resection through a systemic review and meta-analysis.Relevant literature was obtained from three databases including PubMed, Embase and Web of Science. Inclusion and exclusion criteria were set for the screening of articles to be selected for further conduction of systemic review and meta-analysis. The HRs of OS of the sublobar group compared with standard resection group were extracted directly or calculated indirectly from included researches.p = 0.784. Further analysis found no difference in subgroups of synchronous and metachronous (from second metachronous lesion), different population region and dominant sex type.Ten researches published from 2000 to 2017 were included in this study, with 468 and 445 MPLC cases for the standard resection group and sublobar resection group respectively. The result suggested that OS of MPLC patients underwent sublobar resection (segmentectomy or wedge resection for at least one lesion) was comparable with those underwent standard resection approach , with HR 1.07, 95% CI 0.67–1.71, This study may reveal that sublobar resection is acceptable for patients with MPLC at an early stage, because of the equivalent prognosis to the standard resection and better pulmonary function preservation. Further research is needed to validate these findings. It was reported that, as a special type of lung cancer, multiple primary lung cancer (MPLC) occurred on 0.2–20% of all lung cancer cases . In 1975At present, surgery is considered as the first choice of treatments for MPLC patients , 5 as feGuidelines such as NCCN, BTS, CSCO, etc. are used for clinical practice currently for lung cancer. According to the eighth edition of AJCC TNM staging system , tumor nA few researches on ground glass opacity (GGO) suggested that those underwent lung-sparing resection could get better prognosis . And mosThe comprehensive online searches were performed by two independent authors from three databases, including PubMed, Embase and Web of Science from the date of inception to Dec 2018. To achieve maximum sensitivity, we combined the terms “multiple primary lung cancer” or “second primary lung cancer” or “multifocal lung cancer” or “synchronous” or “metachronous” and “Lung Neoplasms/secondary or Lung Neoplasms/therapy” as either keywords or MeSH  terms. 3987 records were obtained from the three databases and 2 more from peers familiar with this field. After duplicate removal and rough screening, 48 articles were left for further assessment of full-text and 10 of them were selected for the qualitative and quantitative synthesis of this study. The screening process of this research followed the PRISMA statement [The inclusion criteria included: a) synchronous and/or metachronous MPLC cases were studied; b) the definition of MPLC criteria should be clearly stated c) two or more lesions were surgically removed in each case d) the study included OS (5-year at least) as the main outcome e) HR and 95%CI of OS according to sublobar and standard resection methods, or related tables or graphs for data extraction were provided.The exclusion criteria were as follows. a) articles including abstracts, letters, case reports, reviews and non-clinical studies; b) articles written in languages other than English; c) studies including cases with primary malignant tumor of other organs; d) those had less than 10 cases within each research group; e) researches with low quality according Newcastle-Ottawa Scaling system.Newcastle-Ottawa Scaling system was adopted for the quality assessment of literature, which was composed of three parts: selection (0–4 stars), comparability (0–2 stars) and outcome (0–3 stars). Articles achieving 6 or more stars were considered of high quality. The quality evaluation of each selected article was carried out by two independent researchers.Ln (HR) and SE (standard error) were used for data combination . The HR I-squared test were used to evaluate the heterogeneity of included studies. And p < 0.10 or I2 > 50% was considered relatively high heterogeneity, in which case random-effect model would be applied. Subgroup analysis and sensitivity analysis were performed to find the source of heterogeneity. Publication bias was shown by funnel plot and examined quantitatively be Begg’s test and Egger’s test. Statistical analysis was carried out with R version 3.5.2  and its meta 4.9–4 package. p < 0.05 was considered statistically significant.Cochran’s Q test and Higgins In total 3987 relative articles were obtained from the three databases and after screening, ten studies published from 2000 to 2017 were included for further analysis , as shown in Fig. The HR and 95% CI of OS of postoperative MPLC cases in sublobar group in comparison with standard group were extracted or calculated from each study. It was noted that the OS of metachronous MPLC cases started from the day of second operation. The heterogeneity of the pooled HR showed that Subgroup analysis was carried out according to the type of MPLC (synchronous or metachronous), population region and dominant sex. As shown in Fig. p = 0.788 > 0.05 and p = 0.874 > 0.05 respectively. These results indicated that there was no significant publication bias in this study.As was displayed in Fig. Multiple primary lung cancer (MPLC) is a special type of lung cancer characterized by more than one cancerous lesions independent of each other and is becoming more common clinically. It may be widely accepted that surgery is the principal treatment for MPLC patients. For these multiple primary lesions, some applied radical resection for all while others used sublobar resection. As the surgical strategy for the treatment of MPLC is divergent, this study aimed to evaluate the difference between these two strategies in terms of prognosis. This study included 913 cases that went through resection of at least two lesions from ten independent researches. And the results suggested that sublobar resection was not an indicator for worse prognosis compared with complete standard resection for all lesions . With medium heterogeneity, subgroup analysis based on different type (synchronous or metachronous), geographic origin and dominant sex was carried out, which showed no statistical difference. No publication bias was detected by either qualitative or quantitative methods.As for single lesion, anatomic pulmonary resection is recommended for patients at early stages , which mThe result of the study showed medium heterogeneity, which limited the value of the combined HR. Though subgroup analysis had been applied, the sources of heterogeneity could not be disclosed completely. Besides, as all studies included were carried out retrospectively and randomization could not be obtained in observational studies, there could be selective bias. Therefore, the main limitation of this study was the relatively low quality of included literature. And the relatively strict inclusion and exclusion criteria led to a small number of included studies. Also, the publication language was limited to English for the improvement of literature quality, which inevitably increased publication bias.Although TNM staging play a crucial role in the prediction of the prognosis of NSCLC patients, we were not able to apply it for analysis in this research due to different versions of staging system adopted in each literature. A few researchers found that the highest pT stage of MPLC was an independent prognostic factor , 20. SomA few researches suggested that stage I NSCLC patients with single lesion receiving segmentectomy tended to have better prognosis compared with wedge resection , 28. As For synchronous MPLC lesions adopting sequential resection, De Leyn et al.  preferreIn conclusion, this study may reveal that sublobar resection is acceptable for patients with MPLC at an early stage, because of the equivalent prognosis as the standard resection and better pulmonary function preservation. However, further research is needed to validate these findings.


To identify the association between night shift work and the risk of various cancers with a comprehensive perspective and to explore sex differences in this association. I2 statistic. Subgroup, metaregression, and sensitivity analyses were performed to explore potential sources of heterogeneity. Contour-enhanced funnel plots and the trim and fill method were used together to analyze bias. Linear dose–response analysis was used to quantitatively estimate the accumulative effect of night shift work on the risk of cancer.  We searched PubMed, Embase, and Web of Science for studies on the effect of night shift work on cancer, including case-control, cohort, and nested case-control studies. We computed risk estimates with 95% confidence intervals (CIs) in a random or fixed effects model and quantified heterogeneity using the P < 0.001; I2 = 76.2%). Night shift work increased the cancer risk in both men  and women . Subgroup analyses showed that night shift work positively increased the risk of breast , prostate , and digestive system  cancers. For every 5 years of night shift work, the cancer risk increased by 3.2% .  Fifty-eight studies were eligible for our meta-analysis, including 5,143,838 participants. In the random effects model, the pooled odds ratio (OR) of cancers was 1.15 (95% CI = 1.08–1.22,  This is the first meta-analysis identifying the positive association between night shift work and the risk of cancer and verifying that there is no sex difference in the effect of night shift work on cancer risk. Cancer risk increases with cumulative years of night shift work. Recent years have witnessed a rise in the number of people working late or night shifts in different employment sectors, such as healthcare, construction, transportation, and food preparation , 2. The There were several postulated causal mechanisms that explain how night shift work multiplies cancer risk. First, melatonin, a marker of circadian rhythms, has a fundamental impact on inhibiting carcinogenesis through antioxidation, regulation of immunity, free radical scavenging, and antiangiogenesis . GeneralPrevious studies have revealed that the circadian timing system differs in the sexes, which is mediated by different neuroendocrine contexts, such as sex hormones and their receptors in SCN , 83, 84.Consequently, we conducted a meta-analysis to investigate this sex difference. We also expanded upon previous meta-analyses by not only evaluating the association between night shift work and a specific cancer but also estimating whether there was a dose–response relationship between night shift work and the risk of multiple cancers.We conducted a comprehensive updated search through May 2018 using PubMed, Embase, and Web of Science databases. Two investigators searched for eligible English articles independently. The search terms were “night shift work” or “rotating shift work” or “night work” or “shift work” and “carcinoma” or “neoplasm” or “tumor” or “cancer”. In addition, we manually reviewed the reference lists of articles for additional relevant studies.Studies were included if they satisfied the following criteria: (i) the research was a case-control study, cohort study, or nested case-control study; (ii) the exposure of interest was night shift work, and the outcome of interest was the risk of any type of cancer; (iii) the study reported adjusted risk estimates  with 95% confidence intervals (CIs) or provided sufficient data to allow calculation. Studies were excluded if they satisfied the following criteria: (i) the study did not provide sufficient data; (ii) the study mentioned recurrent cancer; (iii) when more than one article was based on the same study population, we only included the study with the largest number of cases.Data extraction was conducted independently by two authors for the following items: first author, publication year, study location, study design, number of cases, occupation, quality score, definition of exposure, participant sex, type of cancer, adjusted OR with 95% CI, adjusted covariates, and exposure assessment. As the prevalence of tumor is very low, we considered that ORs equaled RRs or HRs, providing similar risk estimates . AccordiTwo authors performed quality assessment using the Newcastle-Ottawa Quality Assessment Scale (NOS) . The scaI2 for the heterogeneity test was ≤50%, a fixed effects model was adopted to pool ORs; otherwise, a random effects model was selected. To explore potential heterogeneity, we performed subgroup analyses, metaregression analyses, and sensitivity analyses. One subgroup analysis was the classification of work schedules; we used a random effects model to evaluate the effect size for cancer on different work schedules [All statistical analyses were performed using Stata version 12.0 . We preferentially extracted adjusted ORs from original articles to evaluate the association between night shift work and cancer risk. If there were no adjusted ORs for specific subgroup analyses, a number of cases and participants would be extracted to calculate OR. The inverse variance method was used to combine ORs. If chedules . To confchedules .A generalized least-squares trend (GLST) model was used to estimate the overall dose–response relationship of night shift work and the risk of cancer by computing risk estimates for different ordinal levels of night shift work. There were at least three ordinal levels of the exposure category in each study. The midpoint of the upper and lower boundaries of each level was considered the average exposure. The upper boundary of the highest level was considered the same as the adjacent category if it was not provided . We usedP values were two-sided, and statistical significance was set at P ≤ 0.05.Potential publication bias was estimated with the Begg funnel plot. Furthermore, the contour-enhanced funnel plot and the trim and fill method were used together to analyze the cause of bias. All reported The characteristics of the abovementioned studies are summarized in P < 0.001), with high heterogeneity  (shown in online P = 0.003) and women , with high heterogeneity in men  and women  (P = 0.031) but not in women .The random effects model was used to pool the ORs, indicating the relationship between night shift work and risk of multiple cancers. The pooled OR was 1.15  . In cancI2 = 61.6%, P ≤ 0.001) whereas those with low-quality scores did not show this positive relationship and had high heterogeneity . Additionally, increased risk was present in studies with more than four adjusted variables. Studies with fewer than four adjusted variables showed no elevated risk of cancer, with high heterogeneity . We performed metaregression analyses to assess the potential heterogeneity sources . It was found that none of the individual studies could powerfully change the positive result.Sensitivity analysis showed that the pooled ORs were stable and did not identify the origins of heterogeneity. After omitting 19 studies by the leave-one-out analyses, we found a stable positive relationship  between night shift work and the risk of cancer, with low heterogeneity (P < 0.001). For every 5 years of night shift work, the risk of cancer increased by 3.2%   .P = 0.001). After combining the trim and fill method and contour-enhanced funnel plot, the result showed that most of the filled studies were outside the 10% line, which indicated that the previously verified bias was mostly caused by the high heterogeneity, not the publication bias , hematological system cancer , and lung cancer  presented a more reliable conclusion. Previous studies have suggested that a common mechanism might be shared among hormone-dependent cancers including prostate cancer in men and breast and ovarian cancers in women [Tissue-specific functions and output circadian rhythms are related to the different cell-based clock genes in periphery . To exclin women , 92. Melin women , 93, whiOne meta-analysis  analyzinOur subgroup analyses also uncovered other meaningful results. One finding demonstrated that prostate, breast, and digestive system cancers were connected with night shift work whereas night shift work did not raise the risk of cancers of the hematological system, reproductive system, lung, and skin. In addition, Yuan et al.  found thQ and I2 values, we found a significant heterogeneity among the studies included in this article ; therefore, we used a random effects model to decrease the heterogeneity. After subgroup analyses, we found that fixed shift work, digestive system cancer, reproductive system cancer, unclassified occupation, interview data collection, and high-quality studies were related to less heterogeneity, representing more reliable results. However, all P values in the metaregression analyses did not reflect a statistical difference, such that heterogeneity could not be explained by metaregression analysis. One-by-one-omitted sensitivity and leave-one-out analyses showed that the pooled risk estimates were stable and positive, even when 19 studies were omitted, until heterogeneity was reduced to 29.8%. Although we did not find an obvious source of heterogeneity, the specific subgroup analyses, such as a more uniform definition of work schedules, unclassified occupation based on population, more detailed interviews, and high-quality studies, could decrease the potential heterogeneity.Through analyzing Theoretical biological mechanisms for the positive relationship between night shift work and cancer risk are complex. First, night shift work and LAN could disturb the normal synchrony with the day–night rhythm and sleeping and diet patterns and bring about circadian disruption, which could suppress the secretion of melatonin . MelatonTo the best of our knowledge, this meta-analysis is the first and most comprehensive of its kind to identify the association between night shift work and risk of cancer from the perspective of diverse cancers and by sex. There were several strengths in our meta-analysis. First, we enrolled a large number of articles, even using strict inclusion criteria. The massive study population could enhance statistical power and ensure more accurate risk estimation. Second, a linear dose–response analysis was used to quantify the association between accumulative years of night shift work and cancer risk. Third, we classified work schedules and found that rotating shift work could increase cancer risk whereas fixed shift work could not. The classification of work schedules could decrease clinical heterogeneity to make the results more reliable. Fourth, 34 of 57 studies were carried out among the general population, such that the pooled OR could be better extended to the entire population. Our meta-analysis also had several limitations. First, a significantly high heterogeneity was discovered. We observed significant variability in the study design, risk estimates, study population, definition of night shift work, and exposure assessment. Each of these aspects may generate heterogeneity. Even though many statistical methods were used, we still had trouble finding an obvious source of potential heterogeneity; therefore, the conclusions reached in our meta-analysis should be interpreted with caution. Second, the lack of a consistent definition of night shift work may lead to a certain degree of misclassification and result in a dilution of the pooled OR . Third, In conclusion, our meta-analysis identified a positive relationship between night shift work and cancer risk, using a comprehensive perspective of common cancers. We revealed that the risk of cancer increases cumulatively by 3.2% for every 5 years of night shift work. Moreover, we found no difference between men and women in the association between night shift work and the risk of cancer. Overall, on the grounds that public health is adversely affected by night shift work and its prevalence is on the rise, it is indispensable to develop shift work schedules with the aim of reducing cancer risk. Our meta-analysis does not merely increase public awareness, it also supports the recommendation for regular cancer screening among night shift workers.


Hence, we decided to perform a meta-analysis to evaluate the association.Methods/main results: Two investigators search databases systematically and independently. Odds ratios and 95% confidence intervals were used to pool the effect size. Four articles with 618 cases and 932 controls in total were included in our meta-analysis.Conclusions: MPO -463G > A was not associated with CKD susceptibility in recessive model and homozygote comparison. MPO -463G > A was associated with increased risk of CKD in allelic comparison, heterozygote comparison and dominant model, however, the results lacked stability. Owing to insufficient data, the association between MPO -463G > A and CKD cannot be fully confirmed. In contrast, in the setting of sterile inflammation, MPO and MPO-derived oxidants are thought to be pathogenic, promoting inflammation and causing tissue damage [Myeloperoxidase (MPO) is an oxidative lysosomal enzyme that is available in polymorphonuclear neutrophils and monocytes. MPO utilizes He damage . Patiente damage .MPO -463G > A (rs2333227) is a single nucleotide polymorphism (SNP) in position -463 of MPO gene’s 5′ upstream region. The -463 G creates a stronger SP1 binding site, which can increase MPO expression than -463 A [Previous studies have shown that MPO -463G > A (rs2333227) might be associated with chronic kidney disease (CKD) susceptibility, but sample sizes of those studies are relatively small. Hence, we conducted a meta-analysis to evaluate the association.2.2.1.Without any limitation, two investigators used the following terms to search databases systematically and independently: ’myeloperoxidase or MPO’ and ‘failure or dialysis or injury or ESRD or nephropathy or chronic kidney disease or CKD or end-stage’ and ‘renal or kidney’ and ‘polymorphisms or polymorphism’. PubMed, Embase, Cochrane Library, clinicaltrials.gov, and CNKI databases were searched up to July 26, 2017. We also searched the references of related reviews and studies manually.2.2.Inclusion criteria of this meta-analysis: (1) case-control study about the association between MPO -463G > A polymorphism and human chronic kidney disease; (2) enough genotype data. Exclusion criteria of this meta-analysis: (1) repetitive study (only the study with the largest population was included); (2) lack of enough genotype data; (3) editorial, comment, and review; (4) Genome Wide Association Study; (5) studies in cell lines. Academic dissertation was also reviewed. We try to get detailed genotype data by emailing the author.On the basis of inclusion and exclusion criteria above, two investigators selected studies independently and the investigators resolved divergence by discussion.2.3.Data were extracted by two investigators independently. The investigators resolve divergence by discussion. The information below were extracted: first author’s name, publication year, nephropathy type, cases’ and controls’ characteristics, control groups’ source, country, ethnicity, sample for detection, genotyping method, Hardy–Weinberg equilibrium, number of cases and controls for each genotype.2.4.On the basis of Newcastle–Ottawa Scale (NOS) , two inv2.5.p < .05. To assess the strength of the association between MPO -463G > A polymorphism and chronic kidney disease susceptibility, OR and 95% CIs were counted. We got pooled ORs grom respective combination of single studies by allelic comparison (A vs. G), dominant model (GA + AA vs. GG), recessive model (AA vs. GG + GA), homozygote comparison (AA vs. GG) and heterozygote comparison (GA vs. GG). Z-test with p values less than .05 means statistical significant level.On the basis of the PRISMA checklists , our met2 index were used to assess heterogeneity [p values was less than .10 and/or I2 index was more than 50%; otherwise, we performed fixed-effects model (Mantel and Haenszel method) [p values of Begg’s test (PB) less than .05, and the p values of Egger’s test (PE) less than .05 means a significant publication bias. We did all statistical analyses by using Stata 12.0 software . Except for specified conditions, two-tailed p < .05 means significant. Further statistics analysis was done in allelic comparison (A vs. G) with XLSTAT 2014.4.04 software (Copyright Addinsoft 1995–2014), in which a kind of logistic regression called Correlated Component Regression was used.Q-test and Iogeneity . The ran method) . To asse method)  plot and3.3.1.In total, we obtained 283 articles from databases  = 0). In 3.2.In 3.3.In any comparison and any subgroup including more than two studies, sensitivity analyses were conducted. Because only two studies were included in DN (diabetic nephropathy), sensitivity analyses could not be done.Supplementary data)In group ORD, ORD plus, overall and overall plus, when study Doi K  was exclSupplementary data)Other results showed stability in sensitivity analyses.  and p values of Egger’s test (PE) were used. We did not find significant publication bias. (Supplementary data)The publication bias was evaluated by using Begg’s funnel plot and Egger’s test in every genetic model. In Begg’s funnel plot and Egger’s test, symmetry of funnel plot, 3.5.Further statistics analysis was done in allelic comparison (A vs. G) by using a kind of logistic regression called Correlated Component Regression (CCR). CCR provides reliable predictions even with near multicollinear data. Near multicollinearity occurs when a large number of correlated predictors and relatively small sample size exists as well as situations involving a relatively small number of correlated predictors . In our 4.In group ORD, ORD plus, overall and overall plus, we found MPO -463G > A was not associated with CKD susceptibility in recessive model (AA vs. GG + GA) and homozygote comparison (AA vs. GG), and the results showed stability in sensitivity analyses and no publication bias.In group ORD, ORD plus, overall and overall plus, we found MPO -463G > A was associated with increased risk of CKD in allelic comparison (A vs. G), heterozygote comparison (GA vs. GG) and dominant model (GA + AA vs. GG), however, the results lacked stability. Mostly, the stability was affected by study Doi K  and BouaIn group DN, we cannot conduct sensitivity analyses and publication bias analyses.Moreover, our meta-analysis has several limitations. To date, only four eligible studies can be found and performed meta-analysis. Due to scanty data, subgroup analyses could not be performed well, and in some subgroups, sensitivity analyses and publication bias analyses could not be done. The controls were shared with each other in some studies, which were counted repeatedly. We might miss unpublished studies or studies written by other languages.In Correlated Component Regression, AUC (area under curve) of cross-validation was 0.479 . AUC＜0.5 might indicate a weak association between MPO -463G > A and CKD susceptibility.In conclusion, our results suggested that: MPO -463G > A was not associated with CKD susceptibility in the recessive model and homozygote comparison. MPO -463G > A was associated with increased risk of CKD in allelic comparison, heterozygote comparison and dominant model, however, the results lacked stability. Owing to insufficient data, the association between MPO -463G > A and CKD cannot be fully confirmed, and the result should be explained carefully. Well-designed study with enough data are needed to perfect the current meta-analysis.


WNT genes including WNT3 are strong candidates for NSCL/P, since they are involved in regulating mid-face development and upper lip fusion. This study tested association of the WNT3 polymorphisms, rs-3809857 G/T and rs9890413 G/A, with the risk of NSCL/P in a population of Iranian infants.Nonsyndromic cleft lip and/or palate (NSCL/P) is the most common orofacial birth defect, often attributed to ethnic and environmental differences. Up to now, linkage analyses and genome-wide association studies have identified several genomic susceptibility regions for NSCL/P. The The allelic and genotypic frequencies for each participant were determined in 113 unrelated Iranian subjects with NSCL/P and 220 control subjects using PCR and restriction fragment length polymorphism (RFLP) methods. A p-value of ≤0.05 was considered statistically significant.WNT3 rs3809857 GT genotype was significantly lower  in the NSCL/P (21.2%) than the control group (30.42%). For the WNT3 rs9890413 G/A polymorphism, neither genotype nor allele frequencies were significantly different between the case and control groups.The WNT3 rs3809857 GT genotype may have a protective effect against NSCL/P in Iranian population.Our results indicated that the  Clefts of the lip and/or palate (CL/P) are one of the most common birth defects which are typically classified in syndromic and more common non-syndromic (NSCL/P) forms IRF6), Msh homeobox homolog 1 (MSX1), transforming growth factor alpha (TGFa), and domain containing 2 (CRISPLD2) as well as the 8q24 and 17q22 loci have been suggested to be associated with the etiology of NSCL/P Up to now, several distinct genetic and environmental risk factors have been identified and confirmed for the NSCL/P WNT gene family plays an important role in murine craniofacial embryogenesis WNT genes has been reported to be associated with NSCL/P, and the haplotype of the WNT genes may explain the etiology of NSCL/P WNT3A, WNT5A, and WNT11 are significantly associated with NSCL/P WNT3A gene, rs752107 and rs3121310, were significantly associated with NSCL/P in a Chinese population WNT10A gene was associated with NSCL/P The WNT3 gene and NSCL/P in Caucasian populations. The WNT3 rs3809857 was highly associated with NSCL/P in a Polish population WNT3 rs142167 and rs9890413 were shown to be associated with NSCL/P in a population of Caucasian ancestry According to our knowledge, a few studies have evaluated the association between the WNT3, rs3809857 and rs-9890413, with the risk of NSCL/P using PCR and Restriction Fragment Length Polymorphism (RFLP) in an Iranian population.The aim of this study was to evaluate the association between the two SNPs of WNT3 SNPs with the risk of NSCL/P in a population of Iranian infants. The allelic and genotypic frequencies were determined in 113 unrelated newborns with NSCL/P and 220 newborns as control subjects without clefting.This case-control study tested the association of the etc), metabolic or neurologic disorders or anomalies of other organ systems. Samples were recruited from Mofid Hospital, a referral pediatric center in Tehran, Iran during 2013–2015. Ethical approval for the study was obtained from the Ethics Committee of the School of Dentistry, Shahid Beheshti University of Medical Sciences. Informed consent was obtained from all parents.A clinical examination was performed to look for dysmorphic features such as lip pits. Cases were excluded from the study if there was evidence of other facial or skeletal malformations . Amplification conditions started with an initial denaturation step of 4 min at 94°C, followed by 30 cycles of denaturation at 94°C (45 s), annealing of 61°C (40 s) and 72°C extension (30 s), ended by a final 72°C extension for 5 min. The PCR products of the rs-3809857 G/T and rs9890413 G/A SNPs were digested with the BamHI and MspI restriction enzymes at 37°C overnight . The PCR products were subjected to 8% polyacrylamide gel electrophoresis and stained with silver nitrate. The primer sequences and the pattern of restriction fragments are shown in Briefly, a total volume of 20 χ2) and Fisher’s exact tests using the SPSS statistical software  were performed to compare genotype and allele frequencies in the study groups. The sample was stratified by gender, and genotype and allele frequencies were compared between the study groups.A p-value of ≤0.05 was considered statistically significant. The Chi square  and 53 (46.9%) were boys and girls, respectively, while among them six had unknown gender (5.3%). Of all 220 healthy controls, 217 had known gender: 111 males (50.5%) and 106 females (48.2%), while three (1.3%) had unknown gender.WNT3 rs-9890413 G/A was in Hardy-Weinberg equilibrium .The distributions of genotype using chi-square showed that in the control group for the WNT3 SNPs are shown in WNT3 rs3809857 GT genotype was significantly lower  in the NSCL/P (21.2%) than the control group (30.42%). For the WNT3 rs9890413 G/A SNP, neither genotype nor allele frequencies were significantly different between the case and control groups . When stratified by gender, the WNT3 rs9890413 did not differ significantly between cases and controls in the male and female sub-groups.When the samples were stratified by gender, the MSX1, FGFR1, FGF8, BMP4 as well as WNT gene family have successfully been associated with the CL/P NSCL/P is a genetically complex disorder caused by the interaction of multiple genetic and environmental factors WNT3 gene polymorphisms (rs3809857 and rs9890413) were investigated to determine if any of them are implicated in the etiology of NSCL/P. Our results showed that the WNT3 rs3809857 was associated with NSCL/P. The WNT3 rs-3809857 GT genotype was significantly lower  in the NSCL/P than the control group. This result means that the individuals with the GT genotype showed approximately a two-fold decreased risk of NSCL/P. However, no association between the WNT3 rs9890413 G/A polymorphism and NSCL/P was found. When stratified by gender, the WNT3 rs9890413 did not differ significantly between cases and controls in the male and female sub-groups. When stratified by gender, the WNT3 rs3809857 GT genotype in the male sub-group was significantly lower in the case than the control group .In the present study, the two WNT genes with the risk of NSCL/P. Our results are consistent with the results of the studies conducted by Mostowska et alet alWNT3 rs9890413 in WNT3 northeast China (MAF= 0.03) is lower than the one in the European American (MAF=0.363), and Caucasian populations (MAF=0.32). This distinction may be attributed to the population heterogeneity between the northeast China and the western countries There are a few studies which evaluated the association of WNT gene family plays a critical role in the development of the lip and ectoderm. The WNT gene family consists of structurally related conserved genes which encode secreted signaling glycoproteins proteins that play a fundamental role in developmental and cell-biological processes WNT/beta-catenin signaling pathway The NSCL/P risk factors differ among populations which confirm the importance of testing putative susceptibility variants in different genetic backgrounds WNT genes during human craniofacial development and to identify possible functional variants and/or haplotypes in these genes that may influence the risk of NSCL/P in different populations Mutations and polymorphisms in components of Wnt signaling in NCL/P are not strong. Genome-wide association studies for CL/P did not confirm any role of the WNT signaling genes in the etiology of NSCL/P. Accordingly, further studies are required to explore the role of WNT3 rs3809857 SNP was associated with the non-syndromic CL/P and the WNT3 GT genotype may have a protective effect against NSCL/P in Iranian population.In conclusion, the