This is first meta-analysis to evaluate cancer risk associated with secondhand smoking across all cancers. A literature search was conducted for articles published before June 2014 on Pubmed, SCOPUS, Cochrane library, and CINAHL, and 40 articles on secondhand smoke and the prevalence of cancer among never smokers were selected for final analysis as per the inclusion criteria. Of the 40 articles, 27 were case-control studies and 13 were prospective cohort studies. With respect to overall cancer risk, odds ratio (OR) involving never smokers with significant exposure to secondhand smoke compared to never smokers without such exposure was 1.163 (95%CI 1.058–1.279). Subgroup meta-analyses by study design showed significant positive associations for both case-control studies and prospective cohort studies . The association was stronger in the case of females , lung cancer , and breast cancer . Secondhand smoking may increase the overall risk of cancer for never smokers, particularly lung and breast cancer, and especially in women. Strict implementation of smoking cessation programs should be encouraged, not only to reduce active smoking but also to limit exposure to secondhand smoke. Cancer is a major cause of death worldwide and has been a leading cause of death in Korea for several decades . With thCigarette smoking is widely known as a risk factor for lung cancer. A previous meta-analysis showed that people with any history of smoking had 5.5 times the risk of developing lung cancer compared to those who never smoked. Moreover, current smokers were at 8.43 times the risk of developing lung cancer when compared to those who never smoked, indicating a strong positive correlation, irrespective of the type of lung cancer or country ,7. SinceDespite the well-known carcinogenic nature of smoking, studies on the indirect damage from exposure to secondhand smoke or, to be more exact, sidestream smoke (which comes directly from the tip of a cigarette), are relatively scarce. For example, some studies investigating the relationship between secondhand smoke and cancer have focused on determining the threshold amount of secondhand smoke that can cause cancer or elucidating the underlying biological mechanism. Additionally, sidestream smoke is known to cause cancer by exerting both genotoxic and carcinogenic effects, and indeed, the by-products of cigarette smoking, including tobacco-specific nitrosamines (TSNAs) such as 4-(methylnitrosamino)-1-(3-pyridyl)-1-butanone (NNK)—the main cause of adenocarcinoma—have been detected in the urine of non-smokers exposed to secondhand smoke ,13. ThatPrevious epidemiological studies have provided conflicting outcomes regarding whether secondhand smoking clearly increases cancer risk; however, sufficient evidence is not available to provide a firm conclusion. Additionally, there exist no published meta-analyses investigating the relationship between secondhand smoke and cancer risk across all cancers. Thus, this study is aimed at investigating the relationship between the risk of developing various cancers and exposure to secondhand smoke by systematically reviewing recent reports and conducting a meta-analysis.A systematic literature search was conducted in databases to retrieve articles pertaining to secondhand smoke and cancer risk. We searched for articles published before June 2014 in the Pubmed, SCOPUS (which includes EMbase and ISI Web of Science), Cochrane library, and CINAHL databases using the keywords “passive smoking,” “passive smoke,” “secondhand smoking,” or “secondhand smoke” combined with “cancer.” Only articles written in English were included. The results were exported to a reference manager (EndNote) file. After excluding duplicated articles, three authors  independently reviewed the articles based on the inclusion and exclusion criteria (see below). We also searched the references of each article to identify any other relevant articles. In the case of discrepancy, another author (J.M.L.) was consulted, and the final decision was determined by mutual discussion.The following inclusion criteria were applied to retrieve the relevant articles: (1) study on the prevalence or incidence of cancer associated with secondhand smoke, (2) primary observational study with case-control or prospective cohort design, (3) human study, (4) study on never-smokers, (5) study on any type of histologically confirmed cancer, except skin cancer or carcinoma in situ, and (6) study with clear definition of secondhand smoke or passive smoking. Articles were excluded from the analysis if they met any of the following criteria: (1) in vivo or in vitro study, (2) review articles, letters, case reports, or meta-analysis, (3) study on cancer mortality and not incidence, (4) study with unclear subject selection criteria or those including previous smokers (5) study on childhood cancer, or (6) study on the same population sample and cancer type by same author.For the meta-analysis, the total number of never smokers  in the cancer and control groups and the number (%) or unadjusted odd ratios (OR) of subjects exposed to secondhand smoking were extracted from the selected articles. Moreover, hazard ratios (HRs) adjusted for several covariates were extracted to determine the additional main outcome, in terms of the longest and highest exposure compared to the lowest one. Data were also collected for the following parameters: cancer type, gender, country of origin, and period of exposure to secondhand smoking.In the case of duplicated populations  was used to assess the methodological quality of the observational studies, including those with the case-control and prospective cohort design. The NOS contains 8 items and is scored from 0 to 9 stars. No definite cut-off has been defined for the NOS that signifies a high-quality study. We used the mean values of the studies selected for the current study: 7.5 stars for prospective cohort studies and 6.6 stars for case-control studies. According to the mean values, >7 stars and >6 stars were used as the NOS cut-off for prospective cohort studies and for case-control studies, respectively, to indicate a high-quality study .p value of <0.05 was considered to indicate significant heterogeneity. Moreover, Higgins 2I was calculated, which allows a determination of the proportion of the observed variance that is real. 2I above 50% or 75% suggests the presence of a moderate or high heterogeneity, respectively [X axis and the sample size or variance of the Y axis. In the absence of publication bias, the obtained plot is symmetric and shaped as an inverted funnel [Statistical analyses were performed using Comprehensive Meta-Analysis Version 2.2.064 . We used the random-effects model using the Dersimonian and Laird method  to estimectively . To checd funnel . Orwin’sd funnel . The Trid funnel ,28.The characteristics of the 40 selected studies are presented in p = 0.002) for overall cancer, albeit with substantial heterogeneity . Among the 40 studies, 27 were case-control studies and 13 were prospective cohort studies. According to the study design, the summary effect of case-control studies was significant, with OR of 1.17  and that of prospective cohort studies was 1.16  (p < 0.001)  . The met< 0.001) .p = 0.006) .The meta-analysis was conducted on recently published studies, which included duplicate cohorts as described in the methods section. However, meta-analysis of 36 studies , contrary to the result for men with a HR of 1.59 (95% CI 0.91–2.77). Further subgroup analyses based on the association of secondhand smoking to different types of cancer yielded OR of 1.24  for 15 studies on breast cancer and OR of 1.25  for 12 studies on lung cancer. Analysis of two studies each on bladder cancer and pancreatic cancer did not yield any statistically significant summary estimates.Subgroup analysis was conducted to account for the different parameters. The summary effect HR for subgroup analysis of the studies on women alone was noteworthy, at 1.25  for case-control studies was 6.6 stars and was 7.5 stars for prospective cohort studies, and accordingly, 26 studies were classified as high-quality studies and 14 studies as low-quality ones. Subgroup analysis according to the confirmed methodological quality showed that the significance of the results was maintained: OR of secondhand smoking for cancer in high-quality studies was 1.15 (95% CI 1.02–1.30) and OR of secondhand smoking for cancer in low-quality studies was 1.19 (95% CI 1.02–1.39). Likewise, subgroup analysis by region (Western countries or Asia) also maintained the significance of the results .Subgroup analysis was performed using all exposure data, classified by the period of exposure to secondhand smoking: lifetime exposure, exposure during adulthood, and exposure during childhood. Fifteen studies were related to the risk of cancer from the exposure to secondhand smoking during childhood, and 28 studies were related to the risk from exposure during adulthood. Analysis of both subgroups did not yield any statistically significant results. Twenty-two studies examined the risk from lifetime exposure, and the OR for summary effect was statistically significant, at 1.14 (95% CI 1.02–1.27). p = 0.785 in Begg and Mazumdar rank correlation). The OR in observed studies using Orwin’s method, however, was 1.05, suggesting that the impact of this bias was probably trivial. Similarly, after the trim and fill adjustment using Duval and Tweedie’s Trim and Fill method, the adjusted risk ratio was 1.05 (95% CI 0.95–1.15), which remained fairly unchanged. (Asymmetry existed in the funnel plot for the observed studies indicating the presence of publication bias (changed. .Secondhand smoking is widely regarded as harmful for health and an environmental risk factor in terms of public health. What is intriguing, however, is the lack of a systematic meta-analysis of its effect on the increase of overall cancer risk among humans. Thus, this study was designed as a meta-analysis on 40 epidemiological studies to systematically verify whether exposure to secondhand smoking does indeed increase the risk of cancer. The findings of this study confirmed that exposure to secondhand smoking significantly increases the risk of overall cancer for never smokers. In particular, secondhand smoking significantly increases the risk of breast and lung cancer and cancers in women.N-nitrosamine, polycyclic aromatic hydrocarbons, aldehydes and nickel [TP53 due to miscoding. In addition, they affect tumor genes such as p53 and KRAS, which deregulates the process of cell growth regulation, suppresses apoptosis, and eventually causes unrestrained growth of cells [Tobacco smoking is widely known to be one of the major causes of cancer. Tobacco smoke causes exposure to approximately 7000 kinds of chemicals and 70 kinds of carcinogens such as benzopyrene, chloroethylene, d nickel . Among tof cells ,63.N-nitrosodimethylalanine at concentrations that are 4.5 times and a phenomenal 100 times greater than that in mainstream smoke, respectively [Secondhand smoking can be classified as “mainstream smoke,” which is exhaled by a smoker after inhaling cigarette smoke, or as “sidestream smoke,” which comes directly from the tip of a cigarette. Sidestream smoke is known to contribute to 80% of the smoke in secondhand smoking  and sincectively . In factectively . Furtherectively .Considering these results, it is reasonable to assume that, similar to firsthand smoking, secondhand smoking is also related to the onset of cancer. A previous meta-analysis  showed tTherefore, this study was designed as a meta-analysis of epidemiological studies focusing on the relationship between secondhand smoking and cancer. Only studies including never-smokers were selected in order to enable the evaluation of the effect of secondhand smoking alone. Since the studies varied in design and exhibited heterogeneities with respect to various parameters, the random-effects model was chosen for the meta-analysis. Analysis of the 40 selected studies showed that secondhand smoking increased the risk for all cancers by 16%. Moreover, no significant difference was noted in terms of study design of the studies: both case–control studies and prospective cohort studies showed an increase of 16% in overall cancer risk. The finding that secondhand smoking increased the overall cancer risk by 16% is significant, considering that smoking was responsible for 20.9% of all cancer incidence in Korean adults  and thatSubgroup analyses were further conducted with various parameters since the studies showed heterogeneity in terms of populations and types of cancer. Many studies focused on female subjects alone, since women are more likely than men to be exposed to secondhand smoking, and such studies have shown that the risk of cancer increased by approximately 30%, which portends that secondhand smoking may have a greater effect on women. In addition, analysis based on cancer type demonstrated that secondhand smoking increased the risk of breast cancer by 24% and lung cancer by 25%. Other types of cancers showed no statistically significant results, or only single studies were conducted on some cancer types, whereby the results were insignificant. Statistical significance was maintained in the analysis based on the quality of studies and on the classification of study according to region, i.e., Western and Asian studies. Although sensitivity of individuals may differ with the type of cancer, gender of subjects, and period of exposure to secondhand smoke, it is of vital importance to disseminate that secondhand smoking raises the risk of overall cancer.Trim and Fill method were similar to the existing results, thereby indicating that the bias was only marginal. Fourth, selection and recall bias may be possible in this study because it involves only observational studies. Fifth, since only never smokers were selected for this meta-analysis, there is a possibility of overestimation of the result, because breast cancer and female patients accounted for the majority of the selected study subjects. Sixth, genetic and ethnic susceptibility to cancer was not taken into consideration in this study. For these limitations, our study could not confirm definite conclusion. Further studies are warranted to investigate the relations of secondhand smoke on other cancers and ethnicity.This study has several limitations. First, the included studies showed significant heterogeneities, which necessitated the use of the random effects model, which considers the average effect size as the estimated mean value of the distribution of effect sizes for heterogeneous populations. Second, this study did not account for other risk factors of cancer; this was because the adjusted factors differed across the studies (adjusted covariates of each study are shown in the Despite these limitations, this meta-analysis has certain merits, including the following: extensive data on secondhand smoking were reviewed precisely by strict inclusion criteria; rather than the fixed-effects model, the random-effects model, which considers weights, was used for analysis of all the effect sizes; subgroup analyses in this study included various factors such as study design and types of cancer. Most importantly, this is the first meta-analysis to investigate the effect of secondhand smoking on overall cancer risk.In conclusion, our study results indicate that secondhand smoking increases the risk of cancer, especially in the case of lung and breast cancers and women. In the light of our findings, we believe that history taking in clinical settings should include thorough inquiry regarding exposure to secondhand smoke at the patients’ workplaces and homes. In addition, the finding that secondhand smoking causes cancer in nonsmokers is important not only to the development of public health policies but also in the social and economic context, and it highlights the need for active efforts by medical professionals and government to minimize exposure of the non-smoking public to secondhand smoke.


Objective: Previous studies indicated potential associations between polymorphisms in genes of VEGF/hypoxia/angiogenesis pathway and risk of urogenital carcinomas However, the results were controversial and inconclusive. Here, we conducted an in-depth meta-analysis to investigate the precise associations between polymorphisms in VEGF/hypoxia/angiogenesis related genes and risk of urogenital carcinomas.Methods: We searched PubMed, Web of Science, EMBASE, and Cochrane Library to identify all eligible publications. Pooled odds ratios (ORs) corresponding with the 95% confidence intervals (CIs) were calculated to evaluate their associations. Subgroup analysis was conducted to further ascertain such relationship and investigate sources of heterogeneity.Results: In the end, a total of 96 case-control studies fulfilled the inclusion criteria were enrolled for 12 polymorphisms in 4 VEGF/hypoxia/angiogenesis related genes. The pooled results showed eNOS-rs2070744 polymorphism conferred a significantly increased overall risk of urogenital carcinomas in allele, homozygote, and recessive models, respectively. In addition, eNOS-Intron 4a/b VNTR polymorphism was identified related to an increased risk of urogenital carcinomas in recessive model. And VEGF-rs699947 polymorphism was also identified an increased risk of renal cell carcinoma (RCC) in allelic, heterozygote, dominant, homozygote, and recessive models.Conclusion: To conclude, eNOS-rs2070744 and eNOS-Intron 4a/b VNTR polymorphisms are risk factors for urogenital carcinomas. VEGF-rs699947 polymorphism was also identified as an increased risk factor for renal carcinoma. Urogenital carcinomas, mainly including renal cell carcinoma (RCC), bladder carcinoma (BCa), and prostate carcinoma (PCa), are common malignancies among human neoplasms, with increasing morbidity worldwide in the last 2 decades . The gene set was originally provided via the KEGG signaling database, and encompassed the following 29 genes: ARNT, EIF1, EIF1AX, EIF2B1, EIF2B2, EIF2B3, EIF2B4, EIF2B5, EIF2S1, EIF2S2, EIF2S3, ELAVL1, FLT1, FLT4, HIF1A, HRAS, KDR, eNOS, PIK3CA, PIK3CG, PIK3R1, PLCG1, PRKCA, PRKCB, PTK2, PXN, SHC1, VEGF, and VHL.The gene set of VEGF/hypoxia/angiogenesis pathway was referenced from the Kyoto Encyclopedia of Genes and Genomes (KEGG) website. The VEGF/hypoxia/angiogenesis pathway gene set could be extracted via following URL link  articles has shown the association between polymorphisms in genes of VEGF/hypoxia/angiogenesis pathway and risk of urogenital carcinomas; (2) case-control studies; (3) publications with sufficient genotype data to assess odds ratios (ORs) and 95% confidence intervals (CIs). The exclusion criteria were as follows: (1) case-only studies without a control group, case reports, conference abstracts or reviews; (2) studies without raw data for the genotype; (3) studies with overlapping data.Two authors (JBC and MZ) individually extracted eligible data from each publication based on the inclusion and exclusion criteria. If there is a discrepancy, we reached agreement after discussing with a third author (XBZ). Information collected as follows: first author name, publication year, ethnicity, genotyping methods, source of controls including population-based (PB) or hospital-based (HB), Hardy-Weinberg equilibrium (HWE), cancer type, number of cases, and controls in the VEGF/hypoxia/angiogenesis genotypes. The Newcastle-Ottawa Scale (NOS) was used for assessing the quality of studies. High-quality study required score 7 to 9, and a score less than 7 defined as a low-quality study (Supplementary Table https://www.ncbi.nlm.nih.gov/variation/tools/1000genomes/) was used to extract the LD data, which comprising the polymorphisms in VEGF, HIF-1α and eNOS evaluated in the present study. Briefly, populations enrolled in the project including JPT , YRI , CEU (Utah residents with Northern and Western European ancestry from the CEPH collection), and CHB . The LD was assessed by r2 statistics in these populations using Haploview software.The 1,000 genomes Project database , recessive (BB vs. BA + AA), dominant (BA + BB vs. AA), homozygous (BB vs. AA), and heterozygous (BA vs. AA) models to analyze each variable . Q-statistic as well as According to the inclusion and exclusion criteria, only 12 polymorphisms in four genes of VEGF/hypoxia/angiogenesis pathway were identified . Details of the enrolled studies were listed in Table For polymorphisms in VEGF gene , a total of 49 case-control studies with 8,070 cases and 10,024 controls have met the inclusion criteria. Specifically, there are 918 cases and 1,330 controls in VEGF-rs10434, 3,522 cases and 3,167 controls in VEGF-rs1570360, 3,106 cases and 4,152 controls in VEGF-rs2010963, 3,582 cases and 4,890 controls in VEGF-rs3025039, 3,252 cases and 4,432 controls in VEGF-rs699947, 2,533 cases and 3,550 controls in VEGF-rs833061, respectively. Twenty-two studies of them were performed in Caucasians, 25 studies in Asians, and others were in mixed ethnic groups (including at least one race). Controls of 35 studies were hospital-based (H-B) and 13 studies were population-based (P-B), only one study didn't show whether it was H-B or P-B. Additionally, the distributions of polymorphisms in VEGF for control groups were consistent with HWE, except for these studies listed here , 24 case-control studies with 4,286 cases and 5,009 controls were included in the current work. Specifically, there are 3,777 cases and 4,429 controls in eNOS-rs1799983, 813 cases, and 854 controls in eNOS-rs2070744, 709 cases and 1,111 controls in eNOS-Intron 4a/b VNTR, respectively. Of them, 23 studies were performed in Caucasian, 1 was in mixed ethnic groups. In addition, among these studies, 13 were H-B, and 11 were P-B. All the studies were consistent with HWE. While for HRAS-rs12628 polymorphism, finally, 4 eligible case-control studies comprising 886 cases and 747 controls were included. Of them, 3 studies were performed in Caucasian, 1 in Asian. Of these studies, 2 were H-B, and 2 were P-B. Two studies were not consistent with HWE , respectively. In the stratification analysis by source of control, an increased risk of urogenital neoplasms was also identified for P-B groups in allele, homozygote, and recessive models . Moreover, when the stratification analysis conducted by cancer type , we also identified an increased risk of BCa in allelic, dominant, homozygote and recessive models . An increased risk of PCa in allelic, dominant, homozygote, and recessive models .The pooled results of five included studies had shown eNOS-rs2070744 polymorphism conferred a significantly higher overall risk to urogenital carcinomas in allele, homozygote and recessive models . Subgroups analysis  identified an increased risk of BCa in homozygote models .For eNOS-Intron 4a/b VNTR polymorphism, a total of six eligible case-control studies were included. The final analysis has shown that eNOS-Intron 4a/b VNTR polymorphism was related to an increased risk of urogenital neoplasms in recessive model  .Overall, there was no significant association between eNOS-rs1799983 polymorphism and the risk of urogenital neoplasms. However, subgroups analysis by cancer type revealed an increased risk of BCa in allelic and heterozygote models  .Overall, there was no significant association between HIF1a-rs11549467 polymorphism and the risk of urogenital neoplasms. Subgroups analysis by cancer type revealed an increased risk of PCa in allelic and heterozygote models . Moreover, in the stratification analysis by source of control, an increased risk of urogenital carcinomas for H-B groups was also found .No significant association was uncovered for the association between HRAS-rs12628 polymorphism and urogenital carcinomas risk. However, when conducting the stratification analysis by HWE status, we identified an increased risk of urogenital carcinomas  in dominant and homozygote model for HWE (Y) groups  .Overall, there was no significant association between eNOS-rs1799983 polymorphism and the risk of urogenital neoplasms. Nevertheless, subgroups analysis by cancer type revealed an increased risk of RCC in heterozygote, dominant, and homozygote models . Moreover, in the stratification analysis by source of control, an increased risk of urogenital carcinomas for H-B groups was also found . Subgroups analysis by cancer type  revealed an increased risk of RCC in allelic, dominant, homozygote, and recessive models .No significant association was found between VEGF-rs3025039 polymorphism and the risk of urogenital carcinomas. Nevertheless, subgroup analysis by ethnicity showed an increased risk of urogenital carcinomas in allelic and homozygote model for Asian population . Moreover, when the stratification analysis conducted by cancer type, we also identified an increased risk of RCC in allelic, heterozygote, dominant, homozygote, and recessive models .Overall, there was no significant association between VEGF-rs699947 polymorphism and the risk of urogenital neoplasms. Nonetheless, an increased risk of urogenital carcinomas for Asian populations in allelic, heterozygote, dominant, and homozygote models were found in the stratification analysis by ethnicity , respectively. LD plots for polymorphisms in each gene were presented in Figure Angiogenesis is well-recognized as a key element for sustained tumor growth and a critical factor for tumor metastasis  groups.P-value according to Bonferroni correction [P < 0.05/(12 polymorphisms × 5 models)] was considered as statistically significant). However, we also prefer to identify markers for specific tumors . For these SNPs, PA < 0.05 was considered as statistically significant for cancer type subgroup analysis. Results suggested VEGF-rs699947, VEGF-rs3025039, and VEGF-rs2010963 polymorphisms may be a potential risk factor for RCC. And eNOS-rs2070744, eNOS-Intron 4a/b VNTR, eNOS-rs1799983, and HRAS-rs12628 polymorphisms may be a risk factor for BCa. In addition, eNOS-rs2070744 and HIF1a-rs11549467 polymorphisms may be a risk factor for PCa. Accordingly, these subgroups results suggested these SNPs might be used as potential diagnostic markers for RCC, BCa, and PCa, respectively. There are several important advantages for the current meta-analysis. First and foremost, we implemented a comprehensive database search to identify all eligible studies in the VEGF/hypoxia/angiogenesis pathway, making our meta-analysis more substantial and persuasive. Second, all included studies were assessed by Newcastle Ottawa Scale, aiming to exclude low quality studies and elevate the overall quality. Third, various subgroup analyses based on ethnicity, HWE status, source of controls and cancer type were performed, trying to further stratify the sources of heterogeneity. Fourth, in order to making our analysis more accurate, the recognized formula was performed to adjust the results. In addition, sensitivity analysis was applied to confirm the stability of included studies, and Begg's funnel plot and Egger's test was utilized to analyze publication bias.While for VEGF , HIF1α (rs11549465 and rs11549467), HRAS-rs12628 and eNOS-rs1799983 polymorphisms, our meta-analysis indicated there were no significant associations between them and urogenital neoplasms susceptibility even in subgroup analysis by ethnicity, HWE status, source of controls and cancer type (multiple testing However, several limitations should also be noted in our present meta-analysis. First, for several polymorphisms, particularly when the case number in the studies was small, it may result in an insufficient power for identifying weak association between these polymorphisms and urogenital neoplasms susceptibility. Second, we currently didn't validate these polymorphisms with related clinical consequences. In the follow-up study, we will focus on the functional aspects of how these polymorphisms affect genes expression and ultimately associate with tumorigenesis. Third, all of the included studies were restricted in English, exclusion of other languages studies may increase the publication bias. Fourth, we didn't consider several potential confounding factors in this study, such as the age, gender, smoking habit, drinking status, and environmental factors, etc. This limited us further study the genetic and environmental interaction model. Therefore, the current result should be cautious interpreted.In summary, in conjunction with other studies, current meta-analysis results suggest that eNOS-rs2070744 and eNOS-Intron 4a/b VNTR polymorphisms are associated with elevated risk of urogenital carcinomas. In addition, VEGF-rs699947 polymorphism was also identified as an risk factor for renal carcinoma. Further well-designed case-control studies with large population size are warranted to strengthen our findings.J-BC, MZ, and X-BZ conceived and designed the study, performed the literature search and data extraction. J-BC, MZ, YC, P-HL, Y-WQ, CL, XC, W-BR, and Q-QL performed the meta-analysis and drafted the manuscript. L-FL, M-FC, H-QC, and X-BZ revised the manuscript. H-QC and X-BZ final approval of the version to be submitted.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Raised circulating concentrations of asymmetric dimethylarginine (ADMA), an endogenous inhibitor of nitric oxide synthase (NOS), have been reported in several rheumatic diseases (RDs). However, the strength of this relationship is unclear. Therefore, the aim of this systematic review and meta-analysis was to evaluate the magnitude and the robustness of the association between ADMA concentrations and RDs. We calculated standardized mean differences . Study heterogeneity was evaluated by meta-regressions and sensitivity analyses according to type of RDs, conventional cardiovascular risk factors, inflammatory markers, and type of ADMA assessment methodology. Thirty-seven studies with a total of 2,982 subjects  were included in our meta-analysis. Pooled results showed that ADMA concentrations were significantly higher in patients with RDs than in healthy controls . However, the between-studies heterogeneity was high. Differences in ADMA concentrations between controls and RDs patients were not significantly associated with inflammatory markers, increasing age, lipid concentrations, body mass index, blood pressure, or methodology used to assess ADMA. Furthermore, subgroup analysis showed no difference across RDs. This meta-analysis showed that, in the context of significant between-study heterogeneity, circulating concentrations of ADMA are positively related to RDs. The dysregulation of innate and adaptive immunologic response, with overproduction of pro-inflammatory cytokines, represents the pathogenic hallmark of RDs3. This excess of cardiovascular mortality, only partially predicted by conventional risk factors, has been linked to early endothelial dysfunction and accelerated arterial stiffening4. In fact, peripheral and central endothelial dysfunction, a measure of nitric-oxide (NO) availability and vasodilatory function, has been shown to be more prevalent in RDs than in the general population7.Compared with the general population, patients with RDs suffer from a significantly reduced life expectancy that is mainly due to atherosclerotic cardiovascular disease8, in the occurrence and progression of endothelial dysfunction, arterial stiffening, and atherosclerotic cardiovascular disease9. Moreover, studies in healthy subjects and different patient groups have confirmed the independent prognostic role of basal plasma ADMA concentrations on cardiovascular and all-cause mortality12. Therefore, raised circulating ADMA concentrations might be a surrogate measure of endothelial dysfunction, early atherosclerosis and increased risk of future cardiovascular events.A strong body of clinical and experimental evidence collected in the last decade has convincingly supported a potential pathogenetic role for the accumulation of asymmetric dimethylarginine (ADMA), an endogenous inhibitor of NO-synthase13, have also been observed in psoriatic arthritis (PsA)15, ankylosing spondylitis (AS)20, systemic lupus erythematosus (SLE)21, Sjogren’s syndrome (SSj)22, systemic sclerosis (SSc)29 and Behçet’s disease (BD)34.Increased circulating ADMA concentrations, initially reported in rheumatoid arthritis (RA) patients free of cardiovascular riskHowever, despite being largely investigated, the association between ADMA concentrations and RDs has generally been studied in populations with relatively small sample size. Therefore, we performed a meta-analysis of all relevant studies to define the magnitude and the robustness of the relationship between ADMA concentrations and RDs.A systematic search of publications in the PubMed, Web of Science and Scopus databases, from inception to March 2018, was performed using the following keywords and terms, and their combination: ‘ADMA AND (rheumatic diseases OR rheumatoid arthritis OR psoriatic arthritis OR ankylosing spondylitis OR systemic lupus erythematosus OR systemic sclerosis OR Sjogren’s syndrome OR connective tissue diseases OR vasculitis OR Behçet’s disease)’. In addition, references of the filtered studies were manually checked for potentially missing eligible studies. Two researchers (GLE and AZ) independently screened the abstracts and reviewed the full articles; disagreement between the investigators, when present, was resolved by a third researcher (PP). Inclusion criteria were: (i) assessment of ADMA, (ii) only adult participants, (iii) comparison of healthy subjects with patients with RDs (case-control design), (iv) English language and (vi) full-text publications.A pre-established protocol, including methods for the analysis, was followed: in particular, we investigated the difference in ADMA concentrations stratifying for the following subgroups (i) RD subtype  (ii) connective tissue diseases (CTD) vs non-CTD, (iii) autoimmune conditions  vs conditions with mixed (autoimmune and inflammatory) features  and autoinflammatory conditions , (iv) biological sample (serum vs plasma) and (v) type of laboratory test (e.g. liquid chromatography-LC vs ELISA).Where available, we extracted demographic, anthropometric, clinical and laboratory data from the selected articles and presented them in table format.35. Meta-regression analysis was carried out to investigate potentially influencing factors on ADMA% difference.Quality of studies was assessed with the Newcastle–Ottawa Scale (NOS)23 the mean and standard deviation were inferred from median and IQR as previously reported36, while in two further studies37 the mean and standard deviation were inferred from median and range as previously reported by Hozo et al.38.Standardized mean differences (SMD) were used to create forest plots of continuous data and to assess differences in ADMA concentrations between healthy controls and patients with RDs. A p < 0.05 was considered statistically significant, and 95% confidence intervals (CIs) were reported. In two studies2 statistic was also calculated 40. Statistical heterogeneity was defined as an I2 statistic value ≥ 50% (38). In analyses in which heterogeneity was high, a random-effects model was used.Heterogeneity of SMD across studies was tested with the Q statistic (significance level at p < 0.10). The I42. Duval and Tweedie “trim and fill” procedure was performed to identify and correct for funnel plot asymmetry resulting from publication bias43. The aim of this method is to remove (trim) smaller studies responsible for funnel plot asymmetry, use the trimmed funnel plot to evaluate the true centre of the funnel, then replacing the omitted studies with the new calculated “missing” studies around the centre (filling). This allows to estimate the number of missing studies and recalculates the effect size including the filled studies. Statistical analysis was carried out using MedCalc for Windows, version 15.4 64 bit  and Stata 14 The influence of each single study on effect size was evaluated by sequentially excluding one study at the time, through sensitivity analysis. Publication bias was assessed by means of Begg’s adjusted rank correlation test and Egger’s regression asymmetry test at the p < 0.05 level of significanceThis meta-analysis was performed according to PRISMA statements.A flowchart summarizing the screening process is described in Fig. 57. The characteristics of these studies, published between 2006 and 2018, are summarized in Table After a full-text review of 81 articles, 44 were rejected because they did not meet the inclusion criteria. Thus, 37 studies were included in the meta-analysisA total of 1,860 RDs patients  and 1,122 healthy controls  were evaluated. Overall, the mean age of participants across all studies was 44.6 years both in RDs patients and in controls , random-effects models were used to perform the analysis. Pooled results showed that ADMA concentrations were significantly higher in patients with RDs .The forest plot for ADMA concentrations in RDs patients and controls is shown in Fig. Sensitivity analysis showed that the effect size was not modified when any single study was in turn removed  and Egger’s tests (p < 0.001) evidenced a significant publication bias, we applied the trim-and-fill method to correct the results. Twelve potential missing studies were added on the left side of the funnel plot to ensure symmetry Fig. . The adjTo explore possible sources of heterogeneity, we investigated differences in SMD across RDs Table . Extreme2 = 96.0%, p < 0.0001) were higher than that in participants with no-CTD disease  although the difference was not significant by meta-regression analysis   when compared with RDs patients with mixed (autoimmune and autoinflammatory) diseases  or with autoinflammatory diseases . Similarly, heterogeneity was not influenced by disease sub-groups.As reported in in Fig. We then investigated C-reactive protein (CRP), erythrocyte sedimentation rate (ESR), systolic blood pressure (SBP), diastolic blood pressure (DBP), lipid concentrations, body mass index (BMI), age, sex, and disease duration (DD) as possible contributors to between-study variance. However, none of these variables was found significantly related to pooled SMD by meta-regression analysis .2 = 94.3%, p < 0.0001) was higher than that in studies measuring ADMA in serum  or detection method (LC vs ELISA) may contribute to heterogeneity. The SMD from studies measuring ADMA in plasma  than that of studies measuring ADMA by ELISA , diseases with mixed autoimmune and autoinflammatory features  and autoinflammatory diseases (e.g FMF). However, the difference in autoimmune vs autoinflammatory features did not account for the observed heterogeneity.Moreover, according to Mc Goonagle 69 suggesting that ADMA is causally involved in the pathophysiology of atherosclerosis.A significant association between raised circulating ADMA concentrations and traditional atherosclerotic risk factors  have been reportedInterestingly, in our study, cardiovascular risk factors did not modify the ADMA SMD. These results question the currently accepted hypothesis that raised circulating ADMA concentrations, a potential measure of the presence of co-existing cardiovascular risk factors, may be associated with increased morbidity and mortality for cardiovascular events in RDs. Therefore, prospective studies with larger sample sizes are warranted to further define the significance of the relationship between ADMA and conventional atherosclerotic risk factor and the independent role of ADMA in the development of atherosclerotic disease in RDs.70. However, in our meta-analyses, we found no significant association between ESR, CRP, disease duration and ADMA concentrations: the cross-sectional design and the small sample size of studies included in this meta-analysis may partly explain this finding.Biomarkers of systemic inflammatory burden (such as CRP and ESR) have been shown to be correlated to circulating ADMA concentrations in the general population and in RDsWe also performed two additional pre-specified meta-analyses to evaluate between-group differences according to the analytical techniques used and the biological matrix tested as a potential source of heterogeneity.73. A recent meta-analysis reported plasma ADMA concentrations in healthy subjects ranging from 0.34 to 1.10 μmol/l73. On the contrary, normal serum ADMA concentrations in healthy subjects are less well defined: a recent study reported a range of normality of serum ADMA concentrations of 0.43–0.96 μmol/74.To date, preferred methods for the determination of plasma and serum ADMA concentrations are LC and ELISA. However, it is still a matter of debate whether these two methods are comparable to each other75. We think that this finding might be related to variability in the selection of healthy controls, pre-analytical and analytical issues, potentially accounting for the large heterogeneity of pooled estimates in this meta-analysis.In our subgroup meta-analysis, we found no difference in ADMA SMD on the basis of the analytic method employed or biological matrix tested, suggesting that the large heterogeneity between studies is not explained by techniques variability. However, as reported in Table Finally, we explored the substantial heterogeneity of pooled analysis by evaluating whether specific studies might influence the final results. After excluding each study in turn, the significance of SMD remained stable in every meta-analysis.Moreover, according to Begg’s method and Egger’s test, our meta-analyses carry a significant publication bias. Thus, the trim and fill method was used to adjust for funnel plot asymmetry: in particular, 12 studies were added to balance the funnel plot. The adjusted SMD, even if decreased, remained significant. This indicated that publication bias did not affect the result of the meta-analysis.77.This meta-analysis has some limitations; First, substantial heterogeneity of the included studies might decrease the significance of the results. Second, we only searched for studies written in English, an aspect that may have introduced publication bias. Third, most of the articles did not report the type of anti-inflammatory or anti-rheumatic drugs used and their effects on outcome. This last aspect is of both clinical and experimental relevance, considering that some prospective studies have reported a positive effect in reducing ADMA concentrations of immunosuppressive therapiesStrengths of this systematic review and meta-analysis on the topic of ADMA in RDs, performed on a relatively large and representative population, include the evaluation of ADMA in different disease phenotypes within the ‘rheumatic diseases’ umbrella and extensive analysis of the source of heterogeneity including pathogenetic and clinical features of RDs, analytical techniques and biological matrixes.Moreover, this meta-analysis provided an estimated ‘effect size’ target of 1.27 umol/L for future interventional studies aiming at reducing ADMA in the population of patients with RDs.In summary, results from this meta-analysis show a significant positive association between elevated circulating ADMA concentrations and RDs.The relatively high heterogeneity of the studies included in the meta-analysis, albeit extensively investigated with a number of additional analyses focused on clinical and demographic characteristics, analytical techniques, biological matrixes, and specific studies, requires further research to confirm, or refute, our findings.


TNF-α) promoter region polymorphisms with recurrent pregnancy loss (RPL) risk. However, the results remain con-troversial and incomplete. Hence, we performed a meta-analysis to evaluate the association of the TNF-α -308G>Aand -238G>A polymorphisms with RPL risk.Multiple studies have been carried out examining the association of tumor necrosis factor-α gene and its corresponding 95% confidence interval (CI).TNF-α -308G>A  and nine studies on TNF-α -238G>A , were included in the meta-analysis. Over-all, we found TNF-α -308G>A to be associated with an increase in RPL risk under the homozygote  and the recessive  models. TNF-α -238G>Awas also significantly associated with increased risk of RPL under the allele model . Stratified analysis revealed a more significant association between the TNF-α -308G>A polymorphismand increased RPL risk in Asians under the homozygote , the dominant and the recessive  models,but not in Caucasians. A non-significant association was, however, identified between TNF-α -238G>A and RPL riskbased on ethnicity. Moreover, TNF-α -308G>A and -238G>A polymorphisms were significantly associated with in-creased risk of RPL in high quality studies and polymerase chain reaction-restriction fragment length polymorphism(PCR-RFLP) subgroups.A total of 29 case-control studies, comprising 20 studies on TNF-α -308G>A and -238G>A polymorphisms are associ-ated with an increased risk of RPL.The present meta-analysis demonstrates that  Recurrent pregnancy loss (RPL) is traditionally definedas the occurrence of three or more ≥3) consecutive pregnancylosses; however, the American Society of ReproductiveMedicine (ASRM) has recently redefined RPL astwo or more pregnancy losses , 2. It i consecut(TNF-α) also plays an important role in the implantation, placentation and pregnancy outcome  and -238G>A (rs361525), are most frequently studied in RPL , and Wanfang. The following combination of MeSH terms and keywords was used:  AND (‘‘-308G>A’’ OR ‘‘rs1800629’’ OR ‘‘-238G>A’’ ‘‘rs361525’’) AND  AND . In addition, the reference lists of retrieved articles, reviews and previous meta-analyses were manually screened for additional studies. In the case where more than one article was published by the same author using the same case series, the study with the largest sample size was investigated. Moreover, no restrictions were placed on language, and only published studies with full-text articles were included.This meta-analysis was conducted and reported in accordance with the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) guidelines. All studies published up to December 1, 2017 reporting the association of the TNF-α -308G>A and -238G>A polymorphisms with RPL risk, and iv. Provides sufficient data for calculation of odds ratio (OR) with 95% confidence interval (CI). In addition, the following exclusion criteria were also used: i. Not relevant to TNF-α -308G>A and -238G>A polymorphisms and RPL, ii. The design is based on family or sibling pairs, iii. No usable data reported, iv. The study only involved a case population, v. Animal studies, vi. Duplicated publications, and vii. Abstracts, case-only articles, editorials, and reviews. If studies had partly overlapped subjects, only the one with the largest sample size was included.Studies fulfilling the following selection criteria were included in this meta-analysis: i. Has original and published data, ii. Uses case-control or cohort design, iii. Examines the associations of TNF-α SNP genotype, and deviation from Hardy-Weinberg Equilibrium (HWE) of the control group. Diverse ethnicities were categorized as Asian, African, Latinos and Caucasian.Two investigators independently reviewed full manuscripts of eligible studies, and the relevant data were extracted into predesigned data collection forms. Any discrepancy was resolved by discussion or consensus by involving a third reviewer when required. The following data were collected from each study: first Authors' surname, year of publication, country of origin, ethnicity of the study population, genotyping method, source of control groups , total number of cases and controls as well as numbers of cases and controls for each TNF-α -308G>A and -238G>A polymorphisms with RPL was estimated by ORs and their 95% CIs. Z-test was carried out to evaluate the statistical significance of pooled ORs . The pooled ORs for both polymorphisms were performed under the following five genetic models: allele model (B vs. A), homozygote model (BB vs. AA), heterozygote model (BA vs. AA), dominant model (BA+BB vs. AA) and recessive model (BB vs. BA+AA). The heterogeneity of studies was assessed by using Cochran’s Q-test and the I2 test. A significance level of <0.10 was used to indicate heterogeneity among studies. Moreover, a high value of I2 indicated a higher probability of the existence of heterogeneity . When between-study heterogeneity was found a random-effects model was performed; otherwise, a fixed-effects model (Mantel-Haenszel method) was employed. HWE of genotype distribution in the controls of included studies was conducted using an online program (http://ihg2.helmholtz-muenchen.de/cgi-bin/hw/hwa1.pl), and P<0.05 was considered as significant deviation from HWE. Stratified analyses were performed according to ethnicity, source of controls and study quality (HWE status). To validate the reliability and stability of the results, sensitivity analysis was performed with a single study in the meta-analysis being removed each time to reflect the influence of the individual data set on the pooled OR. The funnel plot was employed to examine publication bias. Egger’s regression analysis was used for re-evaluation of publication bias. The significance of the intercept was determined by the t test suggested by Egger, with P<0.10 considered as representative of statistically significant publication bias. Funnel plots and Egger’s linear regression tests were used to provide a diagnosis of the potential publication bias. In the presence of bias, we utilized the Duval and Tweedie non-parametric ‘‘trim and fill’’ method to adjust results. The statistical analysis for this meta-analysis was performed by using the comprehensive meta-analysis (CMA) version 2.20 software  using two-sided P values.The association of TNF-α -308G>A polymorphism including 3,461 RPL cases and 3,895 controls, and nine case-control studies , PCR-restriction fragment length polymorphism (PCR-RFLP), sequence-specific oligonucleotide (SSO), PCR-sequence-specific primer (PCR-SSP) and direct sequencing. The distribution of genotypes in the controls of five studies , the dominant model  and the recessive model , but not among Caucasian populations. Interestingly, stratified analysis according to genotyping technique revealed a significantly increased risk of RPL in those studies using PCR-RFLP under the homozygote genetic model , the dominant model  and the recessive model . Subgroup analysis of studies with high quality data did not show a significant association between this polymorphism and increased risk of RPL. Summary results of different comparisons are listed in Table 2.TNF-α -238G>A polymorphism and RPL risk. When all the eligible studies were pooled into the meta-analysis, a significant association was found under the allele model  and 15.45% (13.5-17.4%) respectively, showing a lower frequency in Asians  of the TNF-α -308G>A polymorphism under the allele model (A vs. G: PH≤0.001), the heterozygote model (AG vs. GG: PH=0.009) and the dominant model (AA+AG vs. GG: PH≤0.001). Then, we assessed the source of heterogeneity by meta-regression analysis. Surprisingly, ethnicity, glaucoma types, genotyping methods and study quality did not contribute to substantial heterogeneity in the meta-analysis . Therefore, the relationship between this polymorphism and RPL might vary by ethnicity.The subgroup analyses revealed a significant association between TNF-α polymorphisms with RPL. Their results suggested that the TNF-α -308G>A polymorphism was associated with increased RPL risk. Our results are consistent with their meta-analysis. However, their meta-analysis generated contradictory results due to insufficient power because the number of studies was considerably smaller than that needed to achieve robust conclusions. In addition, due to small size, they could not rule out the possibility that publication bias was undetected. More recently, two new epidemiological studies (TNF-α -308G>A polymorphism on RPL risk in Asian populations. Moreover, we found that Li et al. had overlooked seven studies on the TNF-α -308G>A polymorphism and RPL risk in their meta-analysis. Hence, it may significantly affect their total results and RPL results. In the present meta-analysis, by including more nine case-control studies, we found that the TNF-α -308G>A polymorphism was associated with risk of RPL.Recently, a meta-analysis by Li et al.  with 1,4 studies , 34 haveTNF-α -308G>A polymorphism and RPL risk under the homozygote model. After subgroup analysis by ethnicity, the publication bias disappeared in both Caucasians and Asians.Between-study heterogeneity and publication bias are important issues that cannot be ignored in a meta-analysis , 37. In TNF-α -308G>A and -238G>A polymorphisms were not analyzed in combination with other polymorphisms of TNF-α and the effect of gene-gene interactions on RPL development was not undertaken. The relationship between polymorphisms of TNF-α gene and other genetic and environmental risk factors may be highly complicated, and extensive research is still required to ascertain how exactly the TNF-α polymorphisms affect the susceptibility of an individual to RPL.This meta-analysis has a number of limitations that should be noted. First, we could not perform further subgroup analysis by ethnicity among Caucasians and other ethnicities because of the limited number of published studies. Secondly, we strictly followed the inclusion and exclusion criteria to reduce possible selection bias. However, the Egger’s linear regression test showed little publication bias of overall analysis under the homozygote genetic model. Thirdly, the sources of literature searched were from a limited selection of electronic databases, and we failed to retrieve unpublished studies and articles written in other languages and also unpublished studies that might meet the inclusion criteria. Thus, potentially, publication bias might exist even though the funnel plots were found to be symmetrical. Fourthly, subgroup analyses regarding age, number of miscarriage, and other risk factors such as chronic disease were not conducted since the primary literature lacked sufficient data. Finally, the TNF-α -308G>A and -238G>A promoter region polymorphisms with increased risk of RPL. Moreover, our meta-analysis suggests that TNF-α -308G>A is more likely to be associated with the risk of RPL in Asians than Caucasians. Further large, well-designed case-control studies are needed to confirm these findings.This meta-analysis confirmed the association of the


Nonalcoholic fatty liver disease (NAFLD) has gradually become one of the most common chronic liver diseases in the world. More and more evidence shows that low skeletal muscle mass index (SMI) may play a role in the development of NAFLD. Our aim was to quantify the association between SMI, sarcopenia and the presence and severity of NAFLD.We systematically searched English relevant studies from PubMed, Embase, the Web of Science and the Cochrane Library updated to December 20th, 2018. Studies in which SMI was compared between NAFLD cases and controls were included. So were studies concerning the odds ratio (OR) of NAFLD, non-alcoholic steatohepatitis (NASH) and significant fibrosis in sarcopenia patients. Pooled weighted mean differences and ORs were calculated.Of the 1331 retrieved studies, 19 articles were included. SMI level in NAFLD patients was 1.77  lower than that in normal controls. We also found a significantly higher occurrence risk of NAFLD , NASH  and NAFLD-related significant fibrosis  in sarcopenia subjects.SMI level in patients with NAFLD was lower than healthy people, and patients with sarcopenia have higher occurrence risk of NAFLD, as well as its advanced stages including NASH or NAFLD-related significant fibrosis. Further well-designed prospective studies are required to strengthen the arguments.The online version of this article (10.1007/s12072-019-09964-1) contains supplementary material, which is available to authorized users. Nonalcoholic fatty liver disease (NAFLD) has gradually become one of the most common chronic liver diseases in the world, and approximately 25% of the adult population were affected . It is eSkeletal muscle has been considered to be an important endocrine organ that secretes myokines and participates in postprandial glucose utilization, which is a crosstalk between muscle, adipose tissue, liver and other organs. Relative skeletal muscle mass, represented by skeletal muscle mass index (SMI), is the skeletal muscle mass (SMM) divided by height squared or weight. Several cohort and cross-sectional studies have indicated that SMI  is associated with the Incidence rate of NAFLD –8. In adRecently, more and more studies are focusing on the relationship between relative skeletal muscle mass, sarcopenia and NAFLD, but the answer remains controversial. Some studies have reported that patients with sarcopenia have a lower risk of developing NAFLD , 14, whiThis systematic review and meta-analysis were conducted based on the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) Statement (Table S4). We retrieved English relevant studies from PubMed, Embase database, the Web of Science and the Cochrane Library. The latest literature screened was published on December 20th, 2018. The electronic retrieval strategy included the following terms:  AND . Specific search strategies for each database can be found in supplementary materials. Furthermore, the reference lists of key articles were manually and independently reviewed.Studies inclusion criteria were as follows: (1) observational studies or cross-sectional studies; (2) providing a concrete analysis on the odds ratio (OR) that measuring the association between NAFLD、NASH、 NAFLD-related fibrosis and sarcopenia or the mean SMI in NAFLD cases and controls with 95% CI. Studies exclusion criteria were as follows: (1) cell or animal studies, reviews, comments and letters; (2) duplicated studies; (3) research on irrelevant topics; (4) without necessary data or information. The titles and abstracts of the studies were reviewed by two researchers (CC and XS) independently, and which met the inclusion criteria underwent full-text assessment. Discrepancies were resolved by the third reviewer. The information extracted from each selected study was as follows: first author, year of publication, country, study design, measurement of SMM, calculation of SMI, diagnosis of sarcopenia, diagnosis of NAFLD and fibrosis, population characteristics (disease stages and number of cases and controls), adjustment for confounding variables, adjusted OR with 95% CI and SMI with 95% CI.The methodological quality of the enrolled cohort studies was evaluated by the Newcastle–Ottawa Scale (NOS) checklist and determined according to the selection of study groups, comparability of groups and ascertainment of the outcome. The methodological quality of the cross-sectional studies was evaluated by an 11-item checklist, which was known as Agency for Healthcare Research and Quality (AHRQ) (Table S2). If it was defined ‘YES’, the item would be scored ‘one point’; and the score would be a zero if it was defined ‘NO’ or ‘UNCLEAR’. The quality score of the study was graded as follows: low quality defined as 0–3; moderate quality defined as 4–7; high quality defined as 8–11. The supplementary material provides the bias assessment for the studies included.The outcome measure of the meta-analysis was conducted in five aspects. First, we calculated the weighted mean difference (WMD) of SMI between NAFLD patients and normal controls (NC). Subgroup analysis according to gender was performed. SMM measurement methods were divided into two types: bioimpedance analysis (BIA) and dual-energy X-ray absorptiometry (DEXA). We also performed subgroup analysis based on the different SMM measurement methods. Second, we evaluated the strength of association between the risk of NAFLD, NASH and sarcopenia. Further stratified analyses based on race and sample size were performed. Next, we assessed the risk of significant liver fibrosis in NAFLD patients with sarcopenia compared to NAFLD patients without sarcopenia. And subgroup analyses that based on race and whether the diagnosis of NAFLD-related fibrosis was invasive or non-invasive were performed. Relative skeletal muscle mass represented by SMI is the SMM divided by weight × 100%. We analyzed SMI as a continuous variable and calculated the pooled estimates of mean difference between NAFLD patients and NC. To analyze the strength of the association between NAFLD and sarcopenia, we calculated pooled adjusted OR (odds ratio) with 95% CIs.p value < 0.1 was defined as having significant publication bias. And we conducted Trim-and-fill analysis to evaluate the effect of publication bias on the interpretation of the results. Cochran Q test and I2 test were used to assess the heterogeneity between the studies. The low, medium or high heterogeneity was represented by 25%, 50% or 75% of the I2 value, respectively. The random effects model was selected if the p value was less than 0.1 or the I2 value was more than 50%; otherwise, the fixed effects model was chosen. All statistical analyses were performed by Stata (version 12.0).The publication bias was evaluated by constructing a funnel plot of each study’s effect size against the standard error. Funnel plot asymmetry was evaluated by Egger’s test, and A total of 1331 articles were retrieved, of which, 214 were from Pubmed, 540 were from Embase, 556 were from Web of Science and 21 were from Cochrane. After the exclusion of 350 duplicates, 135 reviews and comments, 226 animal studies, 586 irrelevant studies, and 17 articles without concrete data  showed a lower mean SMI than the group of NC , with WMD − 1.77  and statistically significant between-study heterogeneity   . We performed a subgroup analysis based on the different SMM measurement methods   assessed the adjusted OR of NAFLD prevalence between sarcopenia patients and NC . There was no evident publication bias (p = 0.113) . We then performed stratified analyses among Caucasians and Asians. An increased risk of NAFLD was observed to be related to sarcopenia in the Asian population , with significant between-study heterogeneity     indicated a higher risk of NASH prevalence in sarcopenia patients  . Significant publication bias (p = 0.049) was found in this analysis . However, further analysis with trim-and-fill test showed that this publication bias did not impact the estimates .Four studies  showed a higher prevalence of NAFLD-related significant fibrosis in sarcopenia patients compared to NAFLD patients without sarcopenia, with pooled adjusted OR 1.56  , while publication bias did exist according to Egger’s test (p = 0.047) . However, the further trim-and-fill analysis showed that the pooled log OR before and after trimming was 0.472  and 0.433 , showing that the publication bias had little impact on the interpretation of the results . We then performed stratified analyses among Caucasians and Asians. An increased risk of NAFLD-related significant fibrosis was observed in sarcopenia patients in both the Asian population  and the Caucasian population   than in the non-invasively diagnosed subgroup  . These important findings suggested that the lower the SMI one has, the higher the risk of developing NAFLD. Consequently, we showed that sarcopenia subjects having a significantly increased risk of NAFLD . The association was stronger in Asian participants than in Caucasian participants, but the meaning of the results is consistent, which indicated that the race does not affect the overall pooled results. In the robust subgroup with a sample size greater than 2000, the pooled OR was 1.22  with no significant between-study heterogeneity , indicating that when the sample size of the study is larger, the results tend to be more consistent and stable [Our study performed a comprehensive evaluation of the epidemiologic data focusing on the association between SMI, sarcopenia and the presence and severity of NAFLD. We conducted four separate meta-analyses to better quantify the relations. This meta-analysis is hitherto the first study to explore whether the SMI of NAFLD patients differed from that of normal people. Besides, we first offered a pooled estimate and quantitative assessment of the clinical risk of NAFL, NASH and NAFLD-related significant fibrosis in patients with sarcopenia. We demonstrated that, when compared with normal people, NAFLD patients had a lower mean SMI, with WMD − 1.77 (95% CI − 2.39 to − 1.15). The difference turned out to be more obvious in female subgroup than in male subgroup, and the difference in the BIA subgroup was greater than in the DEXA subgroup. However, the data of male and female research, BIA and DEXA research is consistent in terms of trends and meanings, indicating that gender and SMM measurement method does not affect the conclusions of the overall pooled results. As was reported before , after ad stable , 15, 24.d stable . It is wd stable . The staFirst, skeletal muscle plays a crucial role in insulin signaling as a primary tissue responsible for insulin-mediated glucose disposal . A reducSeveral limitations need to be considered when interpreting the results of our work. First, despite our efforts to analyze comprehensively and accurately, some related studies might be omitted. Second, considering that most of the 19 articles we analyzed were in cross-sectional design and only 2 of them were retrospective cohort studies, we could not fully determine the cause-effect relationship between sarcopenia and NAFLD. Third, the definition of SMI varies from study to study. Most studies defined SMI as SMM divided by body weight , 18–23, In conclusion, our meta-analysis indicated that SMI level in patients with NAFLD was lower than healthy people and sarcopenia is associated with NAFLD and particularly the most advanced forms of NASH and fibrosing NAFLD. Restoration of relative skeletal muscle mass may help prevent the onset of NAFLD or its progression, both in normal people and patients with sarcopenia. Clinicians need to bear in mind a potential diagnosis of NAFLD when confronting patients with low relative skeletal muscle mass or sarcopenia. When the diagnosis is made, lifestyle intervention, as the cornerstone management, can be implemented as soon as possible. Since most of the included studies are cross-sectional, the directionality of our results cannot be completely ascertained. Further well-designed prospective studies are required to consolidate these conclusions.Supplementary material 1 (DOC 573 kb)Below is the link to the electronic supplementary material.


Potential risk of thyroid cancer in patients with inflammatory bowel disease has not been well investigated. The aim of the study was to reveal the relationship between history of inflammatory bowel disease and risk of thyroid cancer.First, 1392 patients with inflammatory bowel disease and 1392 controls were included in a case-control study. All patients did not receive immunosuppressive therapy. A multivariate logistic regression analysis was adopted to determine the relationship between history of inflammatory bowel disease and risk of thyroid cancer. Second, a literature search was performed and eight articles were collected. Pooled odds ratios with 95% confidence intervals were reported for relevant risk estimates in fixed or random effect model.P = 0.032). After Bonferroni correction, association of thyroid cancer risk with history of total inflammatory bowel disease or its two subtypes was not found. In the meta-analysis, patients with total inflammatory bowel disease or ulcerative colitis showed an increased risk of thyroid cancer, but patients with Crohn’s disease did not. Furthermore, inflammatory bowel disease patients with immunosuppressive therapy showed an increased risk of the cancer, but patients without immunosuppressive therapy did not have this finding.In the case-control study, thyroid cancer was more common in patients with inflammatory bowel disease than in controls (Risk of thyroid cancer probably elevates in patients with inflammatory bowel disease. Inflammatory bowel disease (particularly ulcerative colitis) itself and use of immunosuppressant might contribute to the development of the cancer. Thyroid cancer (TC) is one kind of endocrine tumors, which originates in the tissue of thyroid gland . PapillaInflammatory bowel disease (IBD) including ulcerative colitis (UC) and Crohn’s disease (CD) is a precancerous disease of colorectal cancer . BesidesTherefore, we conducted a case-control study and a meta-analysis to evaluate the relationship between the risk of TC and the history of IBD and to further explore the mechanism involved.Outpatients in Tianjin Binjiang Hospital between 1991 and 2000 enrolled in a health system database, which contained the huge data of these patients from baseline to 2015. All these outpatients signed the written informed consents and agreed to participate in a series of clinical studies including this study. A total of 1392 IBD patients  were identified in this database according to ICD-9 codes  and served as the IBD group. All these patients did not have any kind of cancer on admission. And, 1392 age- and gender-matched patients with diverticulitis (codes of diverticulitis 62.11 and 562.13) were randomly collected from the database and served as control group.Research data such as demographic information, personal history, medical history, treatment history, and other useful information from baseline to 2015 were collected from the database. Major endpoint was the development of PTC, and all the patients with PTC should be confirmed pathologically. Total follow-up period of these IBD patients were 27,448 person-years.Most of the IBD patients received 5-ASA treatment and other supportive treatment. They did not receive any kind of biologics or immunosuppressant. They did not receive any surgical therapy.History of smoking was defined as smoking (≥ 7 cigarettes per week) for more than 6 months in one’s life. History of drinking was defined as drinking alcohol (≥ 3 times per week) for more than 6 months in one’s life. History of iodized salt intaking was defined as having iodized salt for more than 6 months in one’s life. History of radiation exposure was defined as having radiation exposure (≥ 1 time per month) for more than 6 months in one’s life.t test, and difference of categorical variables was determined using chi-square test. If a P value was less than 0.05, it was considered to be statistically significant. Relationship between the history of IBD and the risk of PTC was determined using multivariate logistic regression analysis. Odds ratios (ORs), 95% confidence intervals (CIs), and P values were reported. Bonferroni correction was adopted for multiple analyses, and Bonferroni corrected P value was 0.004 (0.05/12 variables). If a P value was less than 0.004, it was statistically significant. These analyses were conducted using SPSS 17.0 .Difference of continuous variables was determined using independent sample The study was separately approved by the ethics committee of Tianjin Medical University General Hospital (TMUGH-2016-0657) and Tianjin People’s Hospital  (TJRMYY-2016-087).A structured literature search in MEDLINE/PubMed, EMBASE, Web of Science, and Cochrane Database of Systematic Reviews was conducted between January 1998 and December 2017. Search parameters were “Thyroid Cancer” [Title/Abstract], “Thyroid Neoplasms” [MeSH Terms], “Inflammatory Bowel Disease” [Title/Abstract], “Inflammatory Bowel Diseases” [MeSH Terms], “Ulcerative Colitis” [Title/Abstract], “Colitis, Ulcerative” [MeSH Terms], “Crohn’s Disease” [Title/Abstract], “Crohn Disease” [MeSH Terms] and their combinations.Inclusion criteria were listed as follows: (1) There were IBD group and control group in each study. (2) Number of TC patients in each group was reported. Only English articles were included. Subsequently, references in the included articles were searched. If any reference met the inclusion criteria, it was also adopted.Two trained reviewers independently performed the data selection according to a predefined survey form, which had several contents: first author, year of publication, country, study design, age at diagnosis, gender, period of follow-up, use of immunosuppressant, size of study, and so on. All data were double entered.P values for heterogeneity > 0.05, the results were performed by fixed effects method. If not, the random effects model was adopted. Then, OR and 95%CI were reported. It was statistically significant, if the 95% CI did not include the value one.Our meta-analysis was conducted according to the recommendations from the Cochrane Collaboration and the Quality of Reporting of Meta-analyses guidelines , 15. TheThe study adopted three strategies to ensure the research quality. (1) Funnel plot obtained by Review Manager Software was used to evaluate the publication bias , 17. (2)P < 0.001). More IBD patients had history of smoking, radiation exposure, or benign thyroid nodule . More IBD patients had family history of TC (P < 0.001). Furthermore, the development of PTC was more common in the IBD patients than in the controls (P = 0.032).In Table P = 0.016; OR 5.37, 95%CI 1.78~16.13, P = 0.003; OR 9.22, 95%CI 1.21~70.54, P = 0.031; OR 3.70, 95%CI 1.04~13.26, P = 0.044; OR 4.12, 95%CI 1.12~15.24, P = 0.033). Family history of TC was also related to the elevated risk of the cancer . After Bonferroni correction, only history of radiation exposure and family history of TC reached the Bonferroni-corrected level of statistical significance.In Table As shown in Fig. Major characteristics of the included studies are presented in Table P for heterogeneity = 0.31). After excluding our study, the fixed-effect pooled OR was 1.72 .In Fig. P for heterogeneity = 0.25) .Five studies explored the possible association between the history of UC and the risk of TC, and the fixed-effect pooled OR was 1.62  .Six studies showed the relationship between the history of CD and the risk of TC. The fixed-effect pooled OR was 1.41 . So, included studies showed good homogeneity. Publication bias was an important problem and could not be ignored. As shown in Fig. Quality evaluation of the meta-analysis was conducted and mainly involved three aspects: participant selection, group comparability, and outcome assessment. All included studies received satisfactory scores Table . Major r“Reverse causality” might be a potential limitation in case-control study. But, the studies in our meta-analysis were nested case-control studies with follow-ups of 6000–236,000 person-years. “Recall bias” was another potential problem in retrospective study. But, all included studies obtained their research data from medical records or health data system. The cases of IBD and TC were collected according to ICD-10 system. So, we did not think these issues affected our results.TC was classified into medullary cancer, follicular cancer, undifferentiated cancer, and PTC. The case-control study focused on the PTC patients. But, most of the included studies in the meta-analysis explored the association of IBD with total TC. This limitation might cause bias, though PTC was the most common tissue subtype and accounted for more than 80% of total TC cases. More research should be conducted in different tissue types of TC.In conclusion, there were some defects in the current studies, but their achievements could be the basis for future research. Based on the results of our case-control study and meta-analysis, we suggested that the risk of TC probably elevated in patients with IBD, and this increase was more significant in its subtype UC. Inflammatory response of IBD itself, use of immunosuppressant and other undefined factors contributed to the development of TC. Well-adjusted and population-based studies should be conducted to verify our conclusion.


Pinteraction > 0.05), and remained in sensitivity analyses. No significant associations of NAFLD with CVD  and cancer  mortality were found. In conclusion, NAFLD is a predictor of increased all-cause mortality but not CVD and cancer mortality. These findings have important implications for decision making in public health and clinical practice, and highlight the urgency of developing effective treatments for NAFLD.Whether nonalcoholic fatty liver disease (NAFLD) is associated with an increased risk of mortality remains controversial. The present study aimed to clarify this issue. A systematic search of PubMed and Embase was conducted through October 2018. Studies providing risk estimates of NAFLD and mortality were included. A random-effects model was employed to calculate summary risk estimates. Subgroup analyses were performed to identify potential effect modifiers. Fourteen studies, involving 498501 subjects and 24234 deaths, were included. Patients with NAFLD were found to be at an elevated risk of all-cause mortality compared with those without . The significantly positive association between NAFLD and all-cause mortality could not be modified by age, sex, follow-up duration, and adjustment for body mass index, diabetes, smoking or hypertension (all  Therefore, determining long-term outcomes, including morbidity and mortality, among NAFLD patients has important implications for decision making in public health and clinical practice.Nonalcoholic fatty liver disease (NAFLD) is regarded as the hepatic manifestation of metabolic syndrome, ranging from simple hepatic steatosis to nonalcoholic steatohepatitis (NASH). NAFLD has become a major cause of chronic liver disease worldwide, and is treated as a public health priority. A recent meta-analysis showed that the global prevalence of NAFLD had reached as high as 25.2%3, type 2 diabetes4, and cardiovascular disease (CVD)5. Hence, it is anticipated that NAFLD is a predictor of increased mortality. However, the results from observational studies on NAFLD and mortality remain controversial. Several studies found that patients with NAFLD were at an elevated risk of all-cause mortality compared with those without9, whereas other studies found no association between NAFLD and mortality13. Moreover, a recent large prospective study found that NAFLD was associated with an increased risk of death from all causes, CVD and cancer in women but not in men14. Two recent meta-analyses consistently showed that NAFLD was not associated with all-cause mortality15. However, these meta-analyses included only around one-third of studies currently available. More importantly, these meta-analyses pooled data from multiple reports that originated from the same cohort 18.In addition to its potential to induce cirrhosis and liver cancer, NAFLD is found to be a well-established risk factor for chronic kidney diseaseTo our knowledge, a comprehensive meta-analysis focusing on NAFLD and mortality was not available. Therefore, we performed this study to investigate the associations between NAFLD and mortality from all causes, CVD and cancer.We reported our results in adherence to the Meta-analysis Of Observational Studies in Epidemiology (MOOSE) statement.To identify potentially eligible studies, we performed a systematic search of PubMed and Embase through October 2018, with no restriction. The search strategy applied in these two databases was provided in detail in Table All follow-up studies that reported risk estimates and corresponding confidence intervals (CIs) on the association of NAFLD with mortality from all causes, CVD or cancer were eligible for inclusion. We only considered studies that used either imaging techniques or liver biopsy for diagnosing NAFLD. We did not impose any restrictions on the presence of comorbidity in the study population. Following studies were excluded: (1) studies using NAFLD patients as the reference group; (2) studies conducted in cancer patients undergoing surgical resection; (3) studies conducted in patients undergoing liver transplantation or bariatric surgery; and (4) studies investigating NAFLD and mortality from a certain disease . Based on the above-mentioned eligibility criteria, two investigators (J.J.H. and H.Y.T.) first scrutinized titles and abstracts to exclude apparently irrelevant studies, and then carefully reviewed the full text to further exclude unrelated studies. Any discrepancies regarding study eligibility were settled by discussion.The following information was extracted from eligible studies: first author’s family name, publication date, study location, study population, the source of cohort, mean age, follow-up duration, the number of NAFLD cases, sample size, the number of deaths, the approach to diagnose hepatic steatosis, outcome assessment, the most fully adjusted risk estimates with corresponding 95% CIs, and adjustment variables. One investigator (H.Y.T.) extracted the required data through an electronic spreadsheet, and then another investigator (J.J.H.) checked the data for accuracy. When multiple reports were derived from the same population-based cohort, we extracted data from the report with the longest follow-up duration and the largest sample size. Any discrepancies regarding the results of data extraction were settled by discussion.19. This scale consists of eight items, which are categorized into three aspects . An individual study could earn a maximum of nine stars after evaluating its three aspects. In the present study, high-quality studies were defined as those earning ≥7 stars. Any discrepancies regarding the results of quality assessment were handled by discussion.Two investigators (H.Y.T. and J.J.H.) used the Newcastle-Ottawa quality assessment scale to independently perform the quality assessment for included studies7 and standardized mortality ratio9 were directly treated as an equivalent to HR20. A random-effects model was employed to combine risk estimate from each individual study. The Hedges Q statistic  and the I2 statistic  were employed to qualitatively and quantitatively assess the statistical heterogeneity among studies, respectively. For studies whose authors reported risk estimates by sex21, by the grade of NAFLD22, or by the number of comorbidities7, we combined these stratum data using a fixed-effects model to approximate risk estimates for the main analysis. Similarly, for one study whose authors reported risk estimates separately for NAFLD patients with and without increased levels of liver enzymes13, we combined these stratum data to obtain risk estimates for NAFLD patients.Hazard ratio (HR) was employed as a common measure to evaluate the association between NAFLD and mortality. Liver-related mortality was defined as mortality from any liver disease, including hepatocellular carcinoma (HCC) and cirrhosis. Relative riskPinteraction for the difference between subgroups was calculated through meta-regression. To identify possible sources of heterogeneity and to evaluate the stability of pooled results, we performed sensitivity analyses with the following approaches: using various exclusion criteria, ignoring a single study in turn, and repeating meta-analysis through a fixed-effects model. We employed Begg rank correlation test23 and Egger linear regression test24 to assess publication bias when there were ≥10 studies. We conducted all data analyses through STATA software , and adopted a statistical significance level of P < 0.05 under two-sided test unless otherwise specified.We performed predefined subgroup analyses to investigate whether the observed association of NAFLD with all-cause and CVD mortality was modified by age, sex, study location, the number of NAFLD cases, NAFLD severity, presence of cirrhosis or fibrosis, methods to diagnose NAFLD, quality score, the source of cohort, follow-up duration and adjustment for body mass index, diabetes, smoking, hypertension, or hyperlipidemia/hypercholesterolemia. A 13 and two reports deriving from the Rochester Epidemiology Project25 were included, because their data were used for different analyses . Thus, a total of 14 reports were included , with substantial heterogeneity  . Nevertheless, on the basis of limited studies, we found that NAFLD was significantly associated with an increased risk of all-cause mortality in women  but not in men . The results of sensitivity analyses are shown in Table 6 to 1.40 (95% CI 1.24–1.58)13. Similarly, repeating the analysis with a fixed-effects model did not materially alter the initial results. Moreover, the significant association of NAFLD with all-cause mortality remained after using various exclusion criteria. Of note, the exclusion of four studies with large sample size (>10000)14 yielded a similar risk estimate , but statistical heterogeneity significantly decreased (I2 changed from 80.0% to 25.0%). Both Begg’s test and Egger’s test did not find evidence of publication bias for the association of NAFLD with all-cause mortality (P > 0.05)  Fig. .Table 1S27 for the association of NAFLD with CVD mortality, involving 471849 subjects and more than 5541 deaths. Our study found that NAFLD was not significantly associated with the risk of death from CVD , with moderate heterogeneity   (Table 8 to 1.20 (95% CI 0.98–1.46)21] did not alter the initial finding that there was no significant association between NAFLD and CVD mortality. In addition, the nonsignificant results on NAFLD and CVD mortality remained after using various exclusion criteria . Both using a fixed-effects model  and ignoring a single study in turn did not materially alter the initial results on NAFLD and cancer mortality.The present study included five individual studies8 examined the association of NAFLD with HCC mortality, one6 examined the association of NAFLD with non-HCC mortality , and one8 examined the association of NAFLD with mortality from gastrointestinal malignancy  and nongastrointestinal malignancy . Based on two studies8 on NAFLD and HCC mortality, our meta-analysis found that patients with NAFLD had a 6.27-fold higher risk of death from HCC compared with those without   Fig. .14 for the association between NAFLD and liver-related mortality, involving more than 255 liver-related deaths among 470775 participants. Our meta-analysis found that NAFLD was associated with an increased risk of death from liver disease , with substantial heterogeneity  ; the results showed that there was no significant difference in overall and CVD mortality between patients with and without NASH8. Interestingly, a prospective cohort study, in which the grade of hepatic steatosis was classified into normal, mild, moderate and severe based on liver echogenicity, found that only moderate to severe steatosis was a predictor of increased all-cause mortality32. Based on few studies, our subgroup analysis by NAFLD severity seemed to suggest that only NASH was associated with an increased risk of death from all causes and CVD; however, this finding should be interpreted with much caution due to limited studies included. More studies are definitely needed to clarify this critical issue in the future. Another question is whether the direction and magnitude of the association between NAFLD and mortality can be modified by sex. In the present study, subgroup analysis by sex showed that NAFLD was significantly associated with an increased risk of all-cause mortality in women but not in men, although the risk difference between subgroups was not significant (Pinteraction > 0.05). However, it should be reminded that these results were based on limited studies. Consequently, these results should be interpreted with caution, and need to be confirmed by large prospective cohort studies. If confirmed, identifying the mechanisms of action underlying the aforementioned risk difference between men and women appears to be interesting and challenging.It is noteworthy that two key questions concerning the association of NAFLD with mortality remain unsolved in the present study. NAFLD is a heterogeneous disease that is comprised of a wide spectrum of histological conditions ranging from simple hepatic steatosis to NASH. Targher 5. CVD is the most frequent cause of death among NAFLD patients33. Interestingly, our meta-analysis suggested that NAFLD could not significantly increase the risk of death from CVD. The potential mechanism behind this phenomenon is unclear. A possible explanation may be that NAFLD patients are in a surveillance program, and benefit from an early and valid risk assessment for CVD, thereby decreasing CVD mortality to the level of the general population. Despite a null association between NAFLD and CVD mortality, our meta-analysis revealed a significant association between NAFLD and all-cause mortality. Considering the significant contribution of NAFLD to the burden of liver-related mortality34 as well as well-known hepatic complications among NAFLD patients, we argue that the increased all-cause mortality in NAFLD patients can be attributable to the increased liver-related mortality. Indeed, our meta-analysis found that patients with NAFLD had a higher risk of death from liver disease compared with those without. Nevertheless, our argument needs to be verified by future studies.It is well established that CVD prevalence is significantly associated with NAFLD, and NAFLD patients have a higher risk of developing CVD than the general population14. Notably, two of them recruited more than 100000 participants14. Thus, these large-scale studies would observe more deaths and yield more accurate risk estimates. Consequently, they were given more weight in the meta-analysis . Moreover, larger studies are inclined to report smaller risk estimates and are conducted with more methodological rigor compared with smaller studies35. Indeed, if pooling risk estimates from the aforementioned four studies, we would obtain a smaller risk estimate . In our subgroup analysis by age, we noticed that the statistical heterogeneity within subgroup obviously decreased, indicating that the age of subjects might also contribute to the observed heterogeneity. It has been shown that older adults have a significantly higher prevalence of NAFLD than young adults2. Therefore, the contribution of age to the observed heterogeneity is likely due to an imbalanced distribution of NAFLD prevalence among different age groups.In the present study, we observed substantial heterogeneity across studies for the combined HR of NAFLD and all-cause mortality. As indicated by our sensitivity analyses, four studies with sample size more than 10000 are possible sources of the observed heterogeneity26 conducted in the population with preexisting conditions, which limits the generalizability of our results to some extent. Nonetheless, our sensitivity analysis showed that the initial results on the association of NAFLD with mortality remained after excluding the above-mentioned five studies. Third, although we extracted the fully adjusted risk estimates, we cannot rule out the potential impact of residual confounding on the pooled results, considering that our results were derived from observational studies where residual confounding always exists. Nonetheless, our subgroup analyses suggested that the observed association between NAFLD and overall and CVD mortality could not be modified by adjustment for body mass index, diabetes, smoking, hypertension, or hyperlipidemia/hypercholesterolemia, supporting that NAFLD by itself is a determinant for mortality. Finally, substantial heterogeneity was detected for the association between NAFLD and all-cause mortality. Nonetheless, we have identified sources of the observed heterogeneity through subgroup and sensitivity analyses. Moreover, clinical and methodological heterogeneity presents for all meta-analyses, especially meta-analysis of observational studies.Our study has several limitations. First, we cannot assess the potential effect of dynamic changes in hepatic steatosis over time on the association between NAFLD and mortality, because all eligible studies did not provide data regarding repeat measurements of hepatic steatosis among their participants. As a result, our results might be subject to non-differential misclassification bias, which possibly attenuates the associations of interest. In addition, our results are based on study-level data; thus, we cannot perform an analysis of competing risk for cause-specific mortality to determine its potential impact on the corresponding results. Second, our study included five studiesIn conclusion, NAFLD is associated with an increased risk of all-cause mortality, while there are no significant associations of NAFLD with CVD and cancer mortality. These findings have important implications for decision making in public health and clinical practice, considering that NAFLD is highly prevalent across many countries and can lead to hepatic or extrahepatic complications. Moreover, our findings highlight that developing effective treatments for NAFLD is essential and should be regarded as a public health priority. Future studies should determine whether the observed association between NAFLD and mortality is limited to patients with NASH or can extend into those with simple hepatic steatosis.Supplementary Information