Previous studies have investigated the association of APOE genotype data (APOE E4+ vs. E4-) for subjects with and without GSD. In all the meta-analyses, we used random-effects models to calculate the odds ratios (ORs) as a measure of association as well as the corresponding confidence intervals (CIs).In this paper, we conducted meta-analyses to examine whether APOE E4 is associated with the risk of GSD. A systematic literature search was performed in PubMed, Cochrane Library, EMBASE, and Google Scholar using the following inclusion criteria: 1) Studies on human subjects; 2) subjects in the control group must undergo ultrasound GSD screening, and presence of GSD in the experiment group can be clearly determined, e.g., diagnosis of GSD through ultrasound screening or a previous history of cholecystectomy or cholelithiasis; 3) the studies reported APOE E4 and risk of GSD . No significant associations were observed in subgroup analyses by gender and mean age. We obtained similar insignificant findings if an additive model was used, if subjects who had E2E4 genotype were excluded, or if low-quality studies were excluded.Our literature search found 13 publications with 14 studies, including a total of 1632 GSD patients and 5001 controls, that met the eligibility criteria and were included in the meta-analyses. We did not find a significant association between APOE E4 on GSD risk. Future studies with large sample sizes that control for important confounding/risk factors are needed to validate our findings and to explore other genetic loci that might affect GSD risk.Our meta-analysis found insufficient evidence for the effect of The online version of this article (10.1186/s12881-019-0843-6) contains supplementary material, which is available to authorized users. Gallstone disease (GSD) is one of the most prevalent biliary tract diseases worldwide , affectiThere are two major types of gallstones: cholesterol stones, which mainly consistent of cholesterol monohydrate crystals and precipitates of amorphous calcium bilirubinate, and pigment stones, which mainly contain calcium bilirubinate. The exact pathogenesis of GSD remains to be determined, and efficient strategies for primary prevention and nonsurgical therapies are still under development.ABCG8) [MUPCDH) [APOE) [The etiology of GSD is multifactorial and involves interaction of genetic and environmental factors. Previous research has identified multiple risk factors for the development of GSD, such as age , 5, femaABCG8) , mucin-l[MUPCDH)  and apol) [APOE) .APOE gene is located on chromosome 19. APOE is a major component of very low-density lipoproteins (VLDLs), which is critical for removing excessive blood cholesterol and maintaining normal cholesterol level. Defects in APOE gene in human can lead to familial type III hyperlipoproteinemia (HLP III) showing impaired clearance of chylomicron, VLDL, LDL and increased blood cholesterol [APOE has 3 polymorphic alleles, E2 , E3 , and E4 . The E4 has been found to be implicated in multiples diseases/disorders, such as impaired cognition, late-onset Alzheimer’s Disease, and ischemic cerebrovascular disease [The lesterol . APOE ha disease , 15.APOE in the development of GSD. Research with APOE-deficient mice showed decreased gallstone formation compared to the wild-type mice, suggesting a role of APOE in gallstone formation [APOE in GSD formation are inconsistent. For example, a positive association was found between APOE E4 genotype and cholesterol crystals in bile, fast cholesterol crystallization in gallbladder bile and a higher cholesterol content in gallstones [Human and mouse model studies have been conducted to examine the role of ormation . Howeverllstones , 17. Howllstones , 19. Morllstones .APOE polymorphisms and GSD risk, with inconsistent conclusions. To the best of our knowledge, two meta-analyses have been conducted to address the relationship between APOE and GSD risk [APOE genotype with GSD risk, we performed this updated meta-analysis which adapted more stringent criteria regarding inclusion of eligible studies and included most recent publications.Previous studies also examined the relationship between GSD risk , 22. TheAs our study used a systematic review and meta-analysis, ethical approval of this study and informed consent statement are not required.APOE genotype data (APOE E4+ vs. E4−) for subjects with and without GSD. We chose the one with a larger sample size if multiple studies used overlapping data.The following inclusion criteria were used to determine study eligibility: 1) Studies on human subjects; 2) subjects in the control group must undergo ultrasound GSD screening, and presence of GSD in the experiment group can be clearly determined, e.g., diagnosis of GSD through ultrasound screening or a previous history of cholecystectomy or cholelithiasis; 3) the studies reported Two authors (LL and JY) performed an extensive literature search in PubMed, Cochrane Library, EMBASE and Google Scholar for papers published before July 18, 2017. The keywords used in the literature search are provided online .APOE genotype data for patients with and without GSD. Any discrepancies were resolved in a group meeting. Quality of the included studies were assessed by two authors (LL and JY) independently using Newcastle–Ottawa scale (NOS) [Two authors (LL and JY) independently extracted the following data from the eligible studies, according to a pre-specified protocol for data extraction: name of the first author, year of publication, participants characteristics including sample size, mean age, distribution of gender, race/country of origin of the participants, diagnosis of GSD, and le (NOS) . ExtractAPOE genotype and GSD risk. We used random-effects models to calculate the ORs and the corresponding 95% confidence intervals (CIs) in all the meta-analyses. Between-study heterogeneity was assessed using I2, and publication bias was evaluated using a funnel plot and Egger’s test.All studies satisfied Hardy–Weinberg equilibrium (HWE) regarding the genotype in the control group. Odds ratios (ORs) were used as a measure of association between the We examined the association by gender, and repeated the analysis by excluding subjects who had E2E4 genotype, and subsequently examined the association using an additive model . Finally, we repeated the analysis by excluding studies of low quality (NOS < 6 stars).p < 0.05 was considered statistically significant. This study was reported according to the PRISMA guidelines [All statistical analyses were performed using Stata 11.2  and SAS version 9.4 . A idelines .No additional data are available.Figure All included publications had been published since 1994. Basic characteristics of the included studies were presented in Table p = 0.338; Fig. p = 0.483). There was no evidence of publication bias in stratified meta-analysis by gender, or by mean age . We found evidence of a publication bias for the meta-analysis using an additive model . We found no association in either the male , or the female subjects .We found no association of APOE E4 with GSD risk . We then assessed the association assuming an additive genetic model, and found no statistically significant dosage effect of APOE E4 allele on the risk of GSD .We then excluded subjects who had E2E4 genotype, and found no significant association of APOE E4 in both the > 50 age group , and in the ≤50 age group . We repeated the analysis by excluding studies of low quality (NOS < 6 stars). Our findings remain essentially unchanged . Finally, five studies (four publications) provided data regarding association of APOE genotype with cholesterol GSD [p = 0.418; I2 = 47.8%, p for heterogeneity = 0.105).We also divided the included studies into two subgroups based on the mean age in the control group: > 50 and ≤ 50 years, and conducted corresponding subgroup analysis. We found no statistically significant association of erol GSD , 32–34. APOE with GSD. In the pooled analysis of 1632 GSD patients and 5001 subjects without GSD, we did not find evidence for significant association of APOE with GSD risk. Similar non-significance was observed in subgroup analysis by gender and mean age, and in other sensitivity analyses.In this paper, we performed a systematic literature search and conducted meta-analyses to examine the association of APOE E2/E3/E4 polymorphisms with GSD risk [p = 0.0003). However, based on the forest plot and the 95% CI (0.97–1.61), this increased risk should not be statistically significant . Similar mistakes  can be found in the other findings throughout the publication. This meta-analysis also included one study which provided genotype frequency for cholesterol gallstone patients and pigment stone patients [APOE E4 and GSD risk in non-Chinese studies. Our updated meta-analysis included publications missed by the previous meta-analysis, and we only retained studies in which presence/absence of GSD can be relatively accurately determined. We did not include two studies despite their relative samples sizes because in one study, the existence of GSD at baseline was determined according to a phone interview of medical history [p = 0.376; I2 = 78.6%, p for heterogeneity< 0.0001; Fig. A previous meta-analysis published in 2012 of 17 studies from 16 publications examined the association of GSD risk . Howeverpatients . Out of  history , and in  history . IncludiAPOE E4 carriers [APOE E4 on GSD risk. For example, the study by Jaime et al. [APOE E4 carriers, compared to non-carriers . Similar findings held when we excluded E2E4 carriers. Selection bias may be underlying the inconsistent findings. In the three studies reporting a positive association, the APOE E4 allele frequency is relatively low in the control group (4–8%) compared to the general population [Out of the 14 studies included for meta-analyses in this paper, only three studies reported a significantly increased risk of GSD in carriers , 28, 35.e et al.  found thpulation .APOE with risk of cholesterol GSD. We got a similar non-significant finding, further supporting that there was no association between cholesterol gallstones and APOE E4 genotype. As in other meta-analyses that only utilized published data, we could not control for potential confounding/risk factors, such as age, sex [APOE on GSD risk might be greatly confounded by such factors, and therefore could influence the validity of any meta-analysis that uses unadjusted results. Therefore, such important confounding factors should be taken into account in future studies on the relationship between APOE and the GSD risk.Our study had several limitations: Although efforts were made in the systematic literature search in an attempt to include as many eligible studies as possible, the pooled sample size is still small. We had to exclude two large studies because the presence/absence of GSD cannot be clearly determined. More studies with larger samples are needed to further validate our findings. The heterogeneity was high for many of the meta-analyses in this study. Additional data of participants for each individual study were limited, and were only available for some of the included studies, making it hard to identify the real sources of heterogeneity. To explore the possible sources of heterogeneity, we performed a random-effects meta-regression analysis by including age, gender and race. However, none of the three variables were statistically significant, and there were 59.4% remaining residual variation due to heterogeneity. Interestingly, we obtained acceptable heterogeneity in a sensitivity analysis including 5 studies on the association of age, sex , ethniciage, sex . The estAPOE E4 with GSD. We found no significant effect of APOE E4 on GSD risk. Future studies with large sample sizes that control for important confounding risk factors are needed to validate our findings and to explore additional genetic loci that might affect GSD risk. Prospective studies that take into account important comorbid factors, such as hypertension, diabetes, and coronary artery disease, are also needed to fully elucidate the relationship between APOE E4 and GSD risk.We performed meta-analyses to examine the association of Additional file 1:Keywords used in the literature search. Combination of keywords used in the literature search of potential publications in PubMed, Cochrane Library, EMBASE and Google Scholar (DOCX 14 kb)


What is the risk of serious infections, opportunistic infections, and cancer in patients with rheumatologic diseases treated with interleukin inhibitors?In this systematic review and meta-analysis of 74 randomized clinical trials comprising 29 214 patients, pooled results suggest that risk of serious infections, opportunistic infections, and cancer is increased in patients with rheumatologic diseases who are treated with interleukin inhibitors compared with placebo.This analysis suggests estimates of risk for infections and cancer associated with the use of interleukin inhibitors that can inform shared decision-making when patients and clinicians are contemplating the use of interleukin inhibitors for rheumatologic diseases. The safety profile of interleukin (IL) inhibitors is not well established.To assess the risk of serious infections, opportunistic infections, and cancer in patients with rheumatologic diseases treated with IL inhibitors.Ovid MEDLINE and Epub Ahead of Print, In-Process & Other Non-Indexed Citations; Ovid MEDLINE Daily; Ovid Embase; Ovid Cochrane Central Register of Controlled Trials; Ovid Cochrane Database of Systematic Reviews; and Scopus were searched .Randomized, placebo-controlled trials that evaluated IL inhibitor therapies in rheumatic diseases and reported safety data were included in the analyses.This systematic review is reported according to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement. Two investigators independently extracted study data and assessed risk of bias and certainty in the evidence. Fixed-effects meta-analysis was conducted to pool odds ratios (ORs) for serious infections, opportunistic infections, and cancers for IL inhibitors vs placebo.The outcomes of interest were the number of serious infections, opportunistic infections, and cancers in individuals receiving IL inhibitor therapies compared with placebo.P < .001, I2 = 0%; high certainty), opportunistic infections , and cancer .In this meta-analysis, 74 studies comprising 29 214 patients  were included. Patients receiving IL inhibitors had a higher risk of serious infections (OR, 1.97; 95% CI, 1.58-2.44; The risk of serious infections, opportunistic infections, and cancer appears to be increased in patients with rheumatologic diseases who are treated with IL inhibitors compared with placebo. This systematic review and meta-analysis examines the risk for serious infections, opportunistic infections, and cancers in individuals receiving interleukin inhibitor therapies for rheumatologic diseases. Interleukin-1 inhibitors , IL-6 inhibitors , IL-12/23 inhibitors , and IL-17 inhibitors  have been approved for clinical use in rheumatologic diseases by the US Food and Drug Administration and by the European Medicines Agency.Interleukins (ILs) are cytokines that play a central role in immune regulation and inflammation by promoting proliferation, activation, migration, and regulation of leukocytes.7 there is a paucity of data regarding the safety profile of these agents. The increased risk of serious and opportunistic infections with biologics, including IL inhibitors, has been a plausible safety concern secondary to blockade of biological pathways leading to immune dysregulation.10 However, the currently available evidence is not sufficient to draw conclusions regarding the safety of IL inhibitors with regard to the risk of serious infections and cancer.11 Establishing the safety data for rare adverse events, such as serious infections and cancer, is challenging because individual clinical trials lack adequate sample size. Previous meta-analyses have successfully identified a significant incidence of rare adverse effects by pooling the data in similar situations in which critical toxic effect signals were missed when looking at individual trials.12 For example, a meta-analysis suggested that treatment with rosiglitazone was associated with a significant increase in the risk of myocardial infarction.13 Similarly, several systematic reviews and meta-analyses have attempted to define the safety of tumor necrosis factor (TNF) inhibitors and supported the risk profile by consistent research evidence.21Although the therapeutic efficacy of these targeted biologics is well established by several clinical trials, systematic reviews, and meta-analyses,Despite widespread use, it is still uncertain to what extent therapy with IL inhibitors may be associated with an increased risk of serious infections and cancer. Therefore, we conducted a systematic review and meta-analysis of published clinical trial data to assess the risk of serious infections, opportunistic infections, and cancer in individuals treated with IL inhibitors for any indicated rheumatologic condition.PRISMA) reporting guideline for systematic reviews and meta-analyses and was conducted following an a priori established protocol.This study followed the Preferred Reporting Items for Systematic Reviews and Meta-analyses . Controlled vocabulary supplemented with keywords was used to search for randomized placebo-controlled trials of IL inhibitor therapy for rheumatic diseases. A detailed search strategy is provided in the eAppendix in the Any randomized, placebo-controlled trials that evaluated IL inhibitor therapies in rheumatic diseases and reported serious infections, opportunistic infections, and/or cancers were included. Inhibitors of the following ILs were considered: anakinra (IL-1), rilonacept (IL-1), canakinumab (IL-1), tocilizumab (IL-6), olokizumab (IL-6), clazakizumab (IL-6), sirukumab (IL-6), sarilumab (IL-6), ustekinumab (IL-12/23), brodalumab (IL-17), secukinumab (IL-17), ixekizumab (IL-17), and guselkumab (IL-23). In studies with multiple intervention arms, data were extracted from the IL inhibitor arm. In case of multiple reports from the same study, the data obtained at the longest follow-up duration were considered.Two of us  screened the titles and abstracts independently; the full texts were screened if the articles met the inclusion criteria. Full text of these selected articles was obtained and evaluated by 2 of us  to confirm eligibility for inclusion. Any discrepancy was resolved via discussion. If there was disagreement between the reviewers, a third investigator (J.B.) was contacted and a decision was made through discussion. Data were recorded in a standardized manner, including the last name of first author, year of publication, disease studied, number of patients in treatment and placebo arms, treatment drug, control drug, dosage of drug, follow-up duration, number of serious infections, number and type of opportunistic infections, and number and type of cancers.22 Certainty of evidence was determined using the GRADE  approach.23The Cochrane Collaboration Risk of Bias Assessment Tool was used to assess for selection bias , performance bias (blinding of participant and personnel), detection bias (blinding of outcome assessment), attrition bias (incomplete data), reporting bias (selective reporting), and other sources of bias.24 Oropharyngeal candidiasis infections were grouped for analysis. Prespecified subgroup analyses were performed to evaluate the risk of serious infections with different IL inhibitor therapies and disease states.The outcomes of interest were the number of serious infections, opportunistic infections, and cancers in individuals receiving IL inhibitor therapies compared with placebo. The serious infections were predefined by study investigators using previously validated measures, as infections resulting in hospitalization, the use of antibiotics, or death. The definition of opportunistic infections was based on a consensus statement by Winthrop et al.25 The continuity correction method suggested by Sweeting et al26 was used to adjust if no events were observed in 1 of the study arms, and studies were excluded from the primary analysis if there were no events in either of the study arms. The continuity correction for the treatment and control arm was 1/(R+1) and R/(R+1) respectively, where R is the ratio of control group to treatment group sizes. Sensitivity analysis was performed without continuity correction. Comprehensive Meta-Analysis, version 3 (Biosta) software was used for all data analysis.27The number of patients who received at least 1 dose of the IL inhibitor represented the denominator of our outcome measure. Fixed-effects meta-analysis was conducted to generate odds ratios (ORs) and 95% CIs. The fixed-effects analysis using the Mantel-Haenszel method was conducted because the studies’ estimates were weighted only according to their estimated variances and, therefore, it is more appropriate for pooling rare events.To estimate the absolute harm increase (number needed to harm [NNH]), we calculated and pooled risk differences from the included studies. The NNH equals the inverse of the pooled risk differences.Meta-regression was performed using the fixed-effects model (method of moments) to explore heterogeneity and evaluate the association of the duration of treatment with the risk of adverse events. We converted all ORs by logarithmic transformation to achieve more symmetrical distributions. The natural logarithm of the OR was the dependent variable, and the duration of follow-up was entered as a covariate. We applied a weighted regression model so that the more precise studies have more influence in the analysis.A cumulative meta-analysis was performed by adding individual studies chronologically, and the results were summarized as each new study was added. The purpose of this technique was to provide a visual presentation for the evolution of evidence over time and determine the point estimates. In addition, leave-1-out analysis was conducted by recalculating the pooled ORs while omitting 1 study in turn to assess the influence of single studies on the overall findings.I2 statistic that expressed the percentage of heterogeneity beyond what is expected by chance. The I2 values greater than 25% were consistent with a low degree of heterogeneity; 50%, moderate degree; and 75%, high degree of heterogeneity.28Heterogeneity was assessed using the P value less than .05 was considered to be statistically significant. If publication bias was detected, the Duval and Tweedie trim-and-fill method was used for adjustment.29Publication bias was assessed using funnel plots, and the Egger regression test with a 2-tailed 109    (eTable 1 in the rtainty) . The resMycobacterium tuberculosis, 2 atypical mycobacterial infections, 1 histoplasmosis, and 6 unspecified.A total of 14 trials reported the incidence of opportunistic infections. The median duration of trial and/or safety follow-up was 54 weeks . These trials included 9998 patients       . The resWe calculated the NNH for all primary outcomes. The NNH was 67 for 1 additional serious infection within a median follow-up of 24 weeks. The NNH for cancer was 250  and, for opportunistic infections, 250 .z value, 3.22; P = .001). However, there were no significant associations of duration of drug use with serious  or opportunistic  infections  but not for opportunistic infections  or cancer   the evidence was derived from randomized clinical trials, (2) the meta-analytic effect estimates were precise, (3) the results were consistent (heterogeneity was low or moderate across studies)  suggests that the risk of serious infections, opportunistic infections, and cancer is increased in patients with rheumatologic diseases who are treated with IL inhibitors compared with placebo. This association is warranted by at least moderate certainty using the GRADE approach. The results are robust; the cumulative meta-analysis suggests that estimates are stable, and subgroup analysis based on drugs and disease state showed consistent results. Subgroups for individual drugs  or diseases  with a limited number of trials suggested that the risk of serious infections may be increased, but results were not statistically significant, likely reflecting the fewer number of events and small sample sizes.112 There have also been several systematic reviews addressing the efficacy and safety of IL-1 inhibition in rheumatoid arthritis with similar findings.114 These studies assessed the infection risk of IL-1 inhibition, but we believe our study is unique in assessing infection risk across all IL inhibitors and is more comprehensive. Our findings are also comparable with the safety profile of TNF inhibitors in rheumatic diseases, suggesting an increased risk of serious infections.115 This finding of an increased number of serious infections is in contrast to a Cochrane database systematic review that compared the adverse effects of biologics  in patients with any disease and reported an increased risk of serious infections that was not statistically significant compared with placebo.116 A later meta-analysis of 106 randomized clinical trials showed that the risk of serious infections was increased in patients with rheumatoid arthritis treated with biologics compared with nonbiologic, traditional disease-modifying antirheumatic drugs, supporting the findings of this analysis.18Several smaller studies of IL inhibitors in individual rheumatic diseases have demonstrated an increased risk of infections, which is consistent with our study.117 Similarly, a French registry (RATIO) collected all cases of nontuberculosis opportunistic infections in patients receiving TNF inhibitors for any indication and reported a 10 times higher incidence of opportunistic infections compared with the general population.118 Another retrospective cohort study involving 236 531 patients reported that the crude incidence and risk of nonviral opportunistic infections among new users of TNF inhibitors compared with those initiating nonbiologic disease-modifying antirheumatic drugs was 2.7 vs 1.7 per 1000 person-years .119 The increased risk of opportunistic infections demonstrated in our study may suggest that the safety profile of IL inhibitors is likely similar to that of TNF inhibitors.The existing evidence for risk of opportunistic infections with IL inhibitor therapy is not yet well established. However, several studies have investigated the risk of opportunistic infections with the use of TNF inhibitors. A meta-analysis involving 32 504 patients with rheumatoid arthritis found that biologic agents  appeared to be associated with a small, but significant, risk of specific opportunistic infections  compared with placebo or disease-modifying antirheumatic drugs.125 Several studies have investigated the risk of cancer with TNF inhibitor therapy, but results are mixed.129 In one meta-analysis, Bongartz et al14 reported an increased risk for cancer  with the use of anti-TNF medications . However, several studies suggested no increased risk of overall cancers with TNF inhibitors.129To our knowledge, the safety data regarding the risk of cancer with IL inhibitor therapy have been limited to individual clinical trials in the absence of combined analysis. The findings in our study suggest that the risk of cancer may be increased with longer IL inhibitor therapy. Although this analysis indicated increased cancer risk with time, it is not conclusive. This increased safety signal should be investigated further by long-term clinical data. Meanwhile, caution must be practiced to adhere to the age-appropriate cancer screening guidelines, and annual screening for skin cancers should also be considered. The studies evaluating the use of IL inhibitors in psoriasis have suggested that risk of cancer is less than or comparable to the general population; the evidence is not sufficient to draw definite conclusions, however, and the evidence may not be generalizable to patients with other rheumatic diseases.130 but to our knowledge, no data are currently available concerning use of IL inhibitors. Moreover, while not accounted for in this analysis, many patients receive other immunosuppressive medications, such as prednisone, in addition to IL inhibitors in clinical practice, which increases the risk of infections and cancers. Similarly, the short duration of follow-up in studies included in this review may not be sufficient to detect the actual cancer risk, which can take years to develop. A more applicable approach would be to assess longer-term data to evaluate cancer risk and length of IL inhibitor therapy, but this protocol was not possible through our approach of using clinical trial data. Furthermore, several clinical trials included in our study had shorter durations of follow-up in the placebo groups compared with the treatment groups. We included the longest available event data for treatment groups that could have biased the results to an overestimation of the true risk, as there was longer follow-up in the treatment groups to detect an adverse event compared with the placebo groups. In addition, we did not consider the differential risk associated with low or high dosages, which may underestimate or overestimate the risk of adverse reactions.Our study has several strengths. The analysis is comprehensive, and the results are robust and consistent across subgroups. We have adjusted for publication bias, and we provided not only the assessment for risk of bias but also evaluated the certainty of evidence using the GRADE approach. The study also has limitations. These results must be interpreted with caution because of factors intrinsic to the analysis of study-level data. This analysis assumes that the risk of infections or cancer is constant throughout the duration of treatment. Previous data have suggested that the risk for infection with TNF inhibitors is highest within the first 90 days of therapy,This systematic review and meta-analysis suggests an increased risk of serious and opportunistic infections with IL inhibitor therapy that may be comparable to those reported for other biologics approved for the treatment of rheumatic diseases. The finding of a possibly increased risk of cancer with long-term IL inhibitor treatment should be taken into consideration and needs to be confirmed by real-world data, such as long-term epidemiologic studies from registries. This analysis provides estimates of toxic effects for infections and cancer associated with the use of IL inhibitors that can inform shared decision-making when patients and clinicians are contemplating the use of IL inhibitors for rheumatologic diseases. As a future study, the comparative safety analysis among individual IL inhibitors should be considered.


Background: Non-alcoholic fatty liver disease (NAFLD) is a serious health problem, but the dose-response relationship between sugar-sweetened beverages (SSBs) and NAFLD remains uncertain. Methods: A systematic review and dose-response meta-analysis were conducted following the PRISMA  guidelines. Review Manager 5.3 and Stata 14.0 were used to combine trials and analyze data. The dose-response meta-analysis was performed by non-linear trend regression. Results: Twelve studies recruiting a total of 35,705 participants were included. The results showed that the consumption of SSBs was associated with 1.39-fold increased odds of NAFLD . The risk of NAFLD rose with an increased consumption of SSBs, while the consumptions of low doses (<1 cup/week), middle doses (1–6 cups/week) and high doses (≥7 cups/week) of SSBs increased the relative risk of NAFLD by 14%, 26% and 53%, respectively . Conclusions: This study demonstrates that consumers of SSBs are at significantly increased risk of NAFLD, and the consumption of SSBs has a dose-dependent effect on the risk of NAFLD. The findings of this study strengthen the evidence base for healthy dietary patterns and are meaningful for the primary prevention of NAFLD. Non-alcoholic fatty liver disease (NAFLD) is the most common chronic hepatic disease, characterized by steatosis and inflammation. NAFLD can progress to non-alcoholic steatohepatitis (NASH), which can lead to the development of liver cirrhosis or cancer in some patients ,2,3. TheSugar-sweetened beverages (SSBs) are defined as beverages containing added caloric sweetener . SSBs, including soft drinks, fruit drinks, carbonated drinks and sport drinks, are one of the most common beverages second only to coffee and tea, and are the main source of artificially added sugar. It has been shown that the intake of SSBs is a risk factor for many medical conditions such as diabetes, obesity and metabolic syndrome ,7,8,9,10This systematic review and meta-analyses were performed following the PRISMA  guidelines.A comprehensive literature search was carried out by two researchers (H.W. Chen and J. Wang). All English and Chinese language articles published to 20 January 2019 were screened from six databases including China National Knowledge Infrastructure (e conducted as follows (English database):  OR (“NAFLD”(Title/Abstract) OR “Fatty Liver, Nonalcoholic”(Title/Abstract) OR “Nonalcoholic Steatohepatitis”(Title/Abstract) OR “Steatohepatitis, Nonalcoholic”(Title/Abstract) OR )) AND (“Carbonated Beverages”(Mesh Terms) OR (“Soft Drinks”(Title/Abstract) OR (“Carbonated drinks”(Title/Abstract) OR “Pop, Soda”(Title/Abstract)) OR “sugary soda”(Title/Abstract))). Chinese database (CNKI) searches: ) AND ((“Tansuanyinliao” (carbonated drinks) OR “hantangqishui” (sugary soda) OR “hantangyinliao” (sugary beverages))).The inclusion criteria were as follows: (1) observational studies ; (2) SSB consumption in relation to nonalcoholic fatty liver disease; (3) adjusted odds ratio (OR), risk ratio (RR), hazard ratios (HR) with coCNKI), PubMed, Web of Science, Medline, Cochrane library and EMBASE. The search details werrresponding 95% confidence intervals (CI) or other information sufficient for their calculation. We excluded the studies if their data or full texts were unavailable. Two researchers independently screened the titles and abstracts of articles. The full texts of all potentially relevant trials were retrieved for further assessment. Language was not limited. Two researchers independently evaluated the quality of each study using the Newcastle-Ottawa Quality Assessment Scale (NOS) . ArticleData were extracted from identified articles by using a standardized extraction form. The important information was collected: (1) author names; (2) country; (3) year of publication; (4) sex of participants; (5) age of participants; (6) study design; (7) numbers of NAFLD cases, non-NAFLD cases and total number of participants; (8) SSB consumption category and SSB category; (9) confounder adjustment.I2 statistic [Review Manager 5.3  and Stata 14.0  were used to combine trials and analyze data. We combined the RRs and their 95% CIs to conduct a new RR and 95% CI. Statistical heterogeneity was assessed by a Chi-squared test and tatistic . FinallyIn our study, all ORs and HR were roughly regarded as RR . Twelve In total, 214 articles were identified through searching six different databases and relevant systematic reviews and meta-analyses. Subsequently, 114 articles were retrieved to Endnote X8 after duplicated articles were removed. The titles and abstracts were screened, and the full texts of potentially relevant 33 articles were retrieved for further assessment. Following to our inclusive and exclusive criteria, only 12 articles were included in our meta-analysis ,25,26,27Characteristics of the included studies are shown in SSB consumption in all studies was measured by the Harvard semi-quantitative food frequency questionnaire (FFQ), or physician-administered food frequency questionnaire.p < 0.00001) (I² = 42%) and the fixed effects model was applied.Twelve studies with 35,705 participants were included in the data analysis for risk of NAFLD. The pooled RR of NAFLD in individuals consuming sugar-sweetened beverages was 1.39 . The difference was statistically significant between groups (0.00001) . There wp = 0.01, p < 0.00001, p = 0.03, respectively). Significant heterogeneity existed in high dose subgroup (I² = 55), and the random effects model was applied in the subgroup meta-analysis.As shown in As shown in p < 0.00001, I² = 42). Sensitivity meta-analysis . Subgroup analysis showed that low doses , middle doses  and high doses  of SSBs increased the relative risk of NAFLD by 14%, 26% and 53%, respectively (p < 0.05). There was no geographical heterogeneity between Asian and non-Asian populations, the consumption of SSBs increased the risk of NAFLD by 51% in Asian populations , and by 38% in non-Asian populations . The results are shown in p < 0.00001) (p < 0.00001)  show0.00001) , and whe0.00001) . These fp for non-linear trend < 0.00001) (The consumption of SSBs (cups/week) was expressed as an ordinal variable . In addi0.00001) . A monotThis study is an updated systematic review and meta-analysis assessing the association between SSBs and risk of NAFLD. The results of this study demonstrated a statistically significant association between SSB consumption and the risk of NAFLD, which is consistent with the results of a previous study . Our metp = 0.01, p < 0.00001, p = 0.03, respectively). Therefore, in order to prevent NAFLD, it is essential to stop the intake of sugary beverages, as even <1 cup/week of sugary beverages will significantly increase the risk of NAFLD by 14% (p = 0.01). Furthermore, the dose-response meta-analysis model revealed a non-linear relationship between SSBs (cups/week) and NAFLD . A monotonically increasing relationship was observed for the consumption of SSBs per week ; However, this dose-response meta-analysis can merely reflect the trends of NAFLD in China, because all relevant data were obtained from Chinese population. [The most important advantage of this meta-analysis is that, to the best of our knowledge, this is the first dose-response meta-analysis evaluating the dose-dependent effects of SSBs on risk of NAFLD. Our results showed that consumptions of low doses (<1 cup/week), middle doses (1–6 cups/week) and high doses (≥7 cups/week) of SSBs significantly increased the relative risk of NAFLD by 14%, 26% and 53%, respectively  are still needed to confirm the dose-response relationship of SSBs and NAFLD.


Background and Aims: Genetic variants in the gene SERPINC1 have been shown to be associated with antithrombin deficiency, which subsequently contributes to the susceptibility to venous thrombosis. However, several other studies have shown conflicting results regarding the association of SERPINC1 gene polymorphisms (rs2227589) with the risk of thrombosis. Hence, in the present study, we conducted a case-control study to further evaluate the association between the variant rs2227589 with antithrombin deficiency in pulmonary embolism (PTE). A pooled systematic analysis was also conducted to evaluate the risk of rs2227589 in venous thromboembolism (VTE) among multiple populations.Methods: This case-control study involved 101 patients and 199 healthy controls. The allele frequency of SERPINC1 variant rs2227589 was analyzed by Sequenom assay. Antithrombin anticoagulant activity was detected using an automatic coagulation analyzer. In addition, a pooled systematic analysis on 10 cohorts consisting of 5,518 patients with VTE and 8,935 controls was performed.Results: In total, 27 (26.7%) PTE subjects were diagnosed as having antithrombin deficiency. Our results showed that antithrombin plasma activity was slightly lower in T allele carriers than that in C allele carriers. However, there was no significant correlation between rs2227589 genotype and antithrombin anticoagulant activity. The recessive model showed that rs2227589 was significantly associated (p = 0.026) with an increased risk {odds ratio [OR]: 2.31, 95% confidence interval [CI] (1.09–4.89)} of Chinese PTE. The pooled systematic analysis of all case-control study and meta-analysis showed that rs2227589 polymorphism was associated with an increased risk of VTE in the additive model  and dominant model .Conclusions: Our study demonstrated that variant rs2227589 is associated with an increased risk of PTE in a Chinese population but no correlation with antithrombin anticoagulant activity. However, pooled systematic analysis of multiple populations showed a significant association between rs2227589 and the risk of VTE in the additive and dominant genetic model. Venous thromboembolism (VTE) is a complex and common cardiovascular disease that includes deep venous thrombosis (DVT), cerebral infarction, and pulmonary embolism (PTE), which is caused by multiple factors. The incidence of VTE is around 0.1–0.2% in Caucasian and American populations . The hosF5, PROC, and SERPINC1) contribute to susceptibility to venous thrombosis . Antithrombin exerts its physiological function by inhibiting procoagulation factors, such as thrombin, factor Xa, factor IIa, and other factors of the blood coagulation system (Genetic variations in coagulation system genes (such as rombosis . Proteinrombosis . SERPINCn system . Antithrn system . There an system .SERPINC1 leading to spontaneous antithrombin polymerization in vivo has been shown to be associated with severe childhood thrombosis (SERPINC1 gene (NG_012462.1:g.5301G > A), was found to be associated with the risk of venous thrombosis in a Dutch population  in 2014 . Most ofBlood samples were collected into trisodium citrate by venopuncture. Plasma samples were obtained by centrifugation at 2,000g for 15 min at 4°C and stored at -80°C until analysis. The antithrombin anticoagulant activity was determined by automatic chromogenic assay on the IL Coagulation systems according to the manufacturer’s instructions . Briefly, the plasma was first incubated with Factor Xa reagent in the presence of an excess of heparin, followed by incubation with a synthetic chromogenic substrate. Quantification of the residual Factor Xa was performed by detecting the absorbance at 405 nm by ACLTOP 700 system (Werfen). The absorbance is inversely proportional to the antithrombin level in the test sample. The activation levels were calculated by the standard curve of each detection. The established normal range of plasma antithrombin levels was 83–128%, which was based on our hospital pathology reference from thousands of previously diagnosed clinical subjects.Genomic DNA was extracted from whole blood using QIAamp DNA Blood kit  according to the manufacturer’s instructions. The candidate single-nucleotide polymorphism (SNP) was genotyped by MassARRAY SNP genotyping platform  among PTE case and control subjects. The following primers for polymerase chain reaction (PCR) amplification were used: forward, 5′-ACGTTGGATGGAAAGGCCTTACCCCAAGAG-3′ and reverse, 5′-ACGTTGGATGTCTCCCTGGTAGTTACAGTC-3′. The genotyping assay extension primers for SNP rs2227589 were 5′-GGAGAGCACTTGAAATGAT-3′. All primers were designed by Sequenom’s MassARRAY Designer software. A specific primer extension was performed to detect single-base polymorphisms in the amplified DNA.SERPINC1 (MeSH Terms) OR rs2227589 (All Fields). Only case-control studies regarding an association between rs2227589 and VTE were considered eligible. The data extracted from the eligible studies included author, publication year, racial and genotype methods, and variants frequency. The control population genotype data of publications followed the Hardy-Weinberg equilibrium.A systematic literature search was performed using PubMed, Science Direct, ISI Web of Knowledge, Google Scholar, and the CNKI Database, and the cutoff date was defined as January 2019. A meta-analysis was performed in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) . Search The eligibility criteria of studies are according to our previous publication . The criz-test was used to evaluate the significance of the pooled OR (post hoc) tests were used for cross-tabulation and multiple comparisons, respectively. P < 0.05 was considered statistically significant.All statistical analyses were performed using software STATA 14.0 and SPSS 19 (IBM). The strength of association was evaluated by pooled odds ratio (OR) and 95% confidence interval (CI). Either a fixed-effects model (Mantel-Haenszel method) or a random-effects model was used for meta-analysis. The ooled OR . Heterogooled OR , and pubooled OR . Nonparaooled OR . SensitiAfter filtering with the inclusion criteria and exclusion criteria, a total of 101 patients were enrolled for plasma antithrombin anticoagulant activity level assay in this study. The general characteristics of PTE patients are presented in SERPINC1 polymorphism (rs2227589) in PTE patients and controls are summarized in SERPINC1 polymorphism (rs2227589) were associated (P = 0.026) with an increased risk of PTE , which was detected in the recessive model but not in the additive and dominant models  showed antithrombin deficiency compared with noncarriers (11/40)  located in the intron may cause antithrombin deficiency. Thus, whether there is an association between rs2227589 and antithrombin levels requires further investigation. The functional consequences of SERPINC1 rs2227589 polymorphism might be directly caused by the regulatory effects of this genetic change or caused by other genetic variants linked to the rs2227589 polymorphism. Nevertheless, our study showed consistency with previous studies in that T allele carriers of VTE had slightly lower antithrombin levels, although this was not significant.Our studies showed that the incidence of antithrombin deficiency is 26.7% in PTE. In our study, the definition of antithrombin deficiency was based on our established antithrombin deficiency normal range (83–128%). The time point of the sample collection, the definition of antithrombin deficiency, and the different selection of patients are the several factors that may affect the frequency of antithrombin deficiency . The hign levels . Howevern levels . Thus, vCYP4V2 (rs13146272), SERPINC1 (rs2227589), and GP6 (rs1613662) (SERPINC1) and VTE  . Our stu1613662) . This fi and VTE . Thus, wIn conclusion, our study showed that variant rs2227589 is associated with an increased risk of PTE in a Chinese population. However, there appears to be no correlation with antithrombin deficiency in PTE. Further validation by employing a larger sample size and multiple-center studies may provide a more reliable conclusion of the association. Pooled systematic analysis showed a significant association between rs2227589 and the risk of VTE in the additive and dominant genetic models of multiple populations. As the contribution of rs2227589 to the risk of VTE may vary with different PTE ethnicities, further investigations into the genetic risk factors and phenotype interactions would be needed to improve the prognosis, prevention, and genetic counseling for populations at high risk of VTE.The project was approved by the ethics committee of the Ethics Committee of Shenzhen People’s Hospital. All procedures performed in studies involving human participants were in accordance with the 1964 Declaration of Helsinki ethical standards. Written informed consent was obtained from the patients for the publication of the patient’s identifiable information.YY and YF prepared the project proposal and study design. YY and QS analyzed all of the genotyping data and conducted the statistical analysis. LX and SL conducted sample collection. QH, MW, and MH conducted antithrombin activity detection experiment. MY assisted with the preparation and revision of the manuscript. All of the authors have read and approved the final manuscript.The study was supported by the National Natural Science Foundation of China (21807072), the Guangdong Provincial Natural Science Foundation (2018A030310674), the Shenzhen Science and Technology Project , and the Guangdong Provincial Science and Technology Project (2017A020214016).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.


The associations of the risk of lung cancer with the vitamin D intake and serum level are controversial. We performed a comprehensive dose-response meta-analysis to evaluate the precise relationships between the above mentioned parameters.We performed a web search of the PubMed, Medline, and Embase databases to identify potential studies that evaluated the relationships between vitamin D intake or serum 25-hydroxyvitamin D (25([OH]D) levels and the risk of lung cancer on December 5, 2017. According to the inclusion and exclusive criteria, 16 studies were included in this meta-analysis. The pooled relative risks (RRs) with 95% confidence intervals (CIs) were used to assess the strength of the associations. A dose-response analysis was conducted to quantitate the relationship between the serum 25(OH)D or vitamin D intake and the risk of lung cancer.P = .018).The pooled RR (highest level vs lowest level) showed that the serum 25(OH)D level was not associated with the risk of lung cancer . A high vitamin D intake was inversely correlated with the lung cancer risk . No significant dose-response relationship was observed between the serum 25(OH)D level and the lung cancer risk. The linearity model of the dose-response analysis indicated that with every 100 IU/day increase in vitamin D intake, the risk of lung cancer decreased by 2.4% (RR = 0.976, 95% CI = 0.957–0.995, A high vitamin D intake provides limited protection against lung cancer carcinogenesis. In 2017, approximately 1.6 million patients died of lung cancer, which is approximately 26% of all cancer-related deaths.,3 Hence, early diagnosis and prevention is important to decrease the morbidity and mortality of this disease.Lung cancer has been the most common and lethal cancer worldwide for several decades. In 2018, lung cancer is estimated to account for 11,2350 new cancer cases  in the United States.Vitamin D is mainly obtained from 2 pathways in humans: synthesis in the skin by exposure to ultraviolet radiation from sunlight and direct intake from dietary food. Vitamin D is hydroxylated to the circulating form—25-hydroxyvitamin D (25[OH]D)—in the liver and transformed into 1,25-hydroxyvitamin D in the kidney. In addition, 25(OH)D has a longer half-life than 1,25(OH)D and is considered an appropriate reflector of the vitamin D levels in serum.–7 Previous studies have reported that the vitamin D level is associated with a decreased risk of different cancers including breast, colorectal, and kidney cancers. However, no association between the vitamin D level and prostate, esophageal, pancreatic, skin, and gastric cancers was reported. Previous meta-analyses have reported an association between high serum 25-hydroxyvitamin D levels and a reduced risk of lung cancer.,17 However, the dose-response relationship between the risk of lung cancer and serum 25(OH)D levels or dietary vitamin intake is unclear. Therefore, this comprehensive dose-response meta-analysis aimed to evaluate the dose-response relationship of the risk of lung cancer with the vitamin D intake and serum 25(OH)D level.Vitamin D takes part in many cell functions including cell apoptosis, differentiation, metastasis, angiogenesis, and proliferation.2,19This meta-analysis was performed according to the latest Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA).2.1A systematical search of the PubMed, Medline, and Embase databases was performed up to December 5, 2017, by 2 reviewers (HW and HJ) using the following search terms: vitamin D or 25-hydroxyvitamin D or 25 hydroxyvitamin D or 25(OH)D) AND lung AND . In addition, the reference lists of the original articles were reviewed, from which other available publications were selected manually. No language restrictions were imposed in the process of searching.2.2The inclusion criteria for the studies were as follows: a case-control or cohort design; reports on the associations of the risk of histologically diagnosed lung cancer with the serum 25(OH)D levels and dietary vitamin D intake; inclusion of relative risk (RR), hazard ration (HR), odds ratio (OR) with 95% confidence interval (CI), or associated data to estimate the association of the risk of lung cancer with the highest versus lowest vitamin D levels; and indication of the number of cases and participants and eligible dose concentration for dose-response analysis. The following studies were excluded from the analysis: studies that did not evaluate the associations between the vitamin D intake and lung cancer risk, and studies that used the serum 1,25(OH)D level as an indicator of the vitamin D level. In the event of duplicate publications, the most complete or most recent publication was used.2.3 All controversial questions were resolved by asking the third author (GW).Data were extracted by 2 independent researchers (HW and QW). The following information were selected according to the criteria listed previously: publication year, the first author's name, country, study design, sample size, vitamin D, or serum 25(OH)D level, measurement method, adjusted variables, risk estimates and 95% CI for evaluating the highest vitamin D levels versus lowest vitamin D levels. We choose the maximally adjusted rations as the only evaluation index for preventing potential confounders when studies reported several multivariable adjusted-effect estimates. When studies did not set groups of lowest dose concentration as reference groups, the EXCEL macro document (RRest9) was used for the reference group transforming, and data was re-calculated according to the instructions. NOS evaluating details including the following 3 aspects awarded a total 9 items: selection of participants and measurement of exposure (4 items), comparability (2 items), and evaluation of methodological quality outcome (3 items). Studies with 7 score or higher score were considered as high quality studies.,23The New Castle–Ottawa quality assessment scale (NOS) system, which has been validated as a comprehensive tool for assessing the quality of observational studies in meta-analysis, were used to assessed study quality.2.4I2 statistic. Cut-off values for I2 were set at 25%, 50%, and 75% for low, moderate, and high degrees of heterogeneity, respectively. When the heterogeneity was <25%, a fixed-effect model was chosen; otherwise, a random-effect model was chosen. When the heterogeneity was significant, sensitivity analysis was performed to assess the robustness of the pooled results by excluding one study at a time. Publication bias was assessed using the Begg rank model and Egger linear model. Subgroup analysis was performed according to the country, mean age, study design, smoking, gender, baseline 25 (OH)D levels, measurement method, pathological type, and study quality. We performed a dose-response meta-analysis by using the correlated natural logs of the RRs or ORs with their standard error across all vitamin D-intake categories. To drive the dose-response curve, the restricted cubic splines with 4 knots at the 5%, 35%, 65%, and 95% percentiles of distribution were used in order to evaluate the potential curvilinear relations. All statistical analyses were performed using Stata 12.0 software .Pooled risk estimates (RR or OR) with 95% CI were used to identify associations between the risk of lung cancer and the vitamin D intake or serum 25(OH)D level. The heterogeneity was evaluated with the This study does not include experiments with animals or humans. Due to the nature of the study, ethical consent was not required.33.1After screening the titles and abstracts of 1254 articles, which were identified from the initial search of online databases, 952 studies were excluded. The eligibility of the remaining 302 studies was assessed by full-text reading. Finally, 16 studies were included in our meta-analysis. The search results and eligible literature selection process are showed in Figure –42 Three studies were conducted in China,,30,39 5 studies were conducted in the United States,,36,38,42 7 studies were performed in Europe,,37,40,41 and 1 study was conducted in Australia. Seven studies reported a mean age of <60 years among participants,,34–39 and 8 studies reported a mean age of >60 years.,32,40,42 Eleven studies were cohort studies,–39,41,42 and 5 studies were case-control studies.–30,36,40 Five studies included individuals who smoked,,36,41,42 and 4 studies included non-smokers.,30,39,41 In terms of the pathological type, 5 studies reported on non-small cell lung cancer (NSCLC),,30,41,42 3 studies reported on small cell lung cancer (SCLC),,41,42 4 studies reported on adenocarcinoma,,29,41,42 and 3 studies reported on squamous carcinoma.,41,42 In addition, 5 studies investigated the association between the vitamin D level and the risk of lung cancer in women,,37,39–41 and 3 studies evaluated this association in men.,36,37 Eight studies considered the mean baseline of 25(OH)D level as >50 nM,,31–33,35 and 4 studies considered this level as <50 nM.,34,36,37 Nine studies detected serum 25(OH)D concentrations using the chemiluminescent immunoassay (CIA) method,,31–34,36 and 3 studies used the radioimmunoassay (RIA) method.,35,37 Eleven studies matched the high score with the NOS scale,,37,39–42 and the remaining studies had a low NOS score.,34,35,38 Eight studies were included in the dose-analysis of the serum 25(OH)D level and lung cancer risk,,32,35–37 and 3 studies were included in the dose-analysis of the vitamin D intake and lung cancer risk.,41,42 The main profiles of the included 16 included articles were summarized in Table –42The 16 studies selected were published between 2006 and 2017 and included 7823 lung cancer patients and 27,2304 control subjects.3.2–30,36 and 8 cohort studies–33,35,37 including 4043 patients and 13,4624 controls were analyzed. Due to the significant heterogeneity  indicated that a random-effect model was applied. The pooled RR for the highest level versus the lowest level was 1.046 D level and lung cancer risk, totally 4 case-control studiesTable P = .349) or non-linearity tests (P = .14) of the dose-response analysis suggested no association between the serum 25(OH)D level and the risk of lung cancer.To determine the relationship between the serum 25(OH)D level and lung cancer risk, a dose-response analysis including 5 cohort studies and 4 case-control studies was performed. As shown in Figure 3.3 and 4 cohort studies,39,41,42 including 3780 patients and 13,7680 controls were analyzed. Since no significant heterogeneity , we used a fixed effect model. The multivariable-adjusted RR of lung cancer for the highest level versus the lowest level of dietary vitamin D intake was 0.854 .The dose-response analysis including 3 cohort studies suggested that a 100 IU/day increase in the vitamin D intake decreased the risk of lung cancer by 2.4% degree D level was performed by omitting one included study at a time and showed stable results in the meta-analysis. Publication bias was evaluated by the Begg test and Egger test. In the analysis of the serum 25(OH)D and the risk of lung cancer, the 4Our meta-analysis indicated that the serum 25(OH)D level was not associated with risk of lung cancer. In addition, the dose-response analysis also showed no association between the serum 25(OH)D level and the risk of lung cancer. The comprehensive meta-analysis, dose-response analysis, and subgroup analysis revealed that an increase in the vitamin D intake was associated with a decrease in the risk of lung cancer. Moreover, the serum 25(OH)D level was not associated with risk of lung cancer in most subgroup analyses except for women, baseline 25(OH)D level >50 nM, SCLC, and squamous carcinoma. colorectal, and kidney, cancers. Mechanistically, vitamin D contributes to the transcription level of cathelicidin antimicrobial peptide genes and the translation of CD14, a co-receptor for identifying bacterial lipopolysaccharides, both of which are important for innate immunity in the lung, and improve host defense. Several cell and animal experiments have found that the active metabolite of 25(OH)D—1,25(OH)2D suppresses angiogenesis and cancer cell growth by inhibiting the response to vascular endothelial growth factor. In addition, 1,25(OH)2D inhibits metastasis and proliferation of lung cancer by prevents other signal pathways including mutations of K-ras and epidermal growth factor receptor and the Wnt/β-catenin pathway.–45 In addition, 1,25(OH)2D also increased the expression of E-cadherin, a glycoprotein that is vital for cell adhesion, and prevented cancer cell metastases. Importantly, 1,25(OH)2D decreased the expression of cyclooxygenase-2 and prostaglandin, and contributed to preventing cancer cell growth and angiogenesis. Although vitamin D theoretically inhibits cancer cell angiogenesis and proliferation, and promotes cancer cell apoptosis, differentiation and metastasis theoretically,–7 it does not stop some cancers occurring including prostate, esophageal, pancreatic, skin, and gastric cancers.–15Studies have reported that an increase in the vitamin D level is associated with a decrease in the risk of different cancers including breast,,17 our meta-analysis included more case-control and cohort studies and suggested no relationship between the serum 25(OH)D level and lung cancer risk. In addition, our dose-response meta-analysis including both linearity and non-linearity tests all confirmed the comprehensive results of our meta-analysis. The results of the non-linearity analysis suggested that the risk of lung cancer decreased when the 25(OH)D concentration was low, up till 60 nM (RR = 0.85), after which the risk increased. Interestingly, our results were consistent with those of a previous dose-response meta-analysis that included only cohort studies, which suggested that a high vitamin D intake cannot provide more protection against lung cancer. Feng et al have found that a high serum 25(OH)D level was not associated with the overall survival of lung cancer. In the subgroup analysis, most results were consistent with those of the comprehensive meta-analysis except the results of the subgroup analysis of women, baseline 25(OH)D levels >50 nM, SCLC, and squamous carcinoma.Although previous meta-analysis suggested that a high concentration of serum 25(OH)D level protects against lung cancer occurring,In the analysis of dietary vitamin D intake and lung cancer risk, we found that the comprehensive results were similar to those of the non-linearity test of dose-response analysis of the serum 25(OH)D levels, which suggested that a high level of vitamin D intake significantly decreases the risk of lung cancer. Therefore, we performed a dose-response meta-analysis of the dietary vitamin D intake and lung cancer risk, which suggested that both results of both the non-linearity and linearity tests were significant. However, the non-linearity tests suggested that when the dietary vitamin D intake exceeded 400 IU (RR = 0.85), it provided limited protection against lung cancer occurring. However, because diet comprises only a portion of the total vitamin D intake, these results may not be conclusive.Our meta-analysis had a few limitations. First, the number of included studies in some subgroup analysis was small, which may make influence on the last conclusions. Second, the original studies did not provide individual data, and the results of our meta-analysis were evaluated by pooled RR and the associated 95% CI, which prevented further detailed analysis and precise results. Hence, our results should be interpreted with caution.In conclusion, our comprehensive meta-analysis indicated no association between a high level of circulating 25(OH)D in serum and the risk of lung cancer. The dose-response analysis of the dietary vitamin D intake indicates that every 100 IU/day intake of vitamin D accounts for a 2.4% decrease in the risk of lung cancer. Therefore, our results suggest that a high level of vitamin D intake provides the limited protection by decreasing the risk of lung cancer. Furthermore, the present meta-analysis suggested that well-designed, large-scale, observational, prospective studies should be conducted in the future to validate the precise relationship between the vitamin D intake, and lung cancer risk.Conceptualization: Guo Wei.Data curation: Hu Jing, Guo Wei.Formal analysis: Guo Wei.Funding acquisition: Guo Wei.Investigation: Hu Jing.Methodology: Qian Wei, Guo Wei.Software: Hu Wei.Validation: Zhou Heng.Writing – original draft: Hu Wei.Writing – review & editing: Guo Wei, Zhou Heng.


Obstructive sleep apnea hypopnea syndrome (OSAHS) means apnea and hypopnea caused by partial or complete obstruction of upper airway collapse during sleep. Gastroesophageal reflux disease (GERD) is believed to be associated with various manifestations in the otorhinolaryngology and has been found to be an additional risk factor for OSAHS.A meta-analysis was performed to identify the association between obstructive sleep apnea hypopnea syndrome and gastroesophageal reflux disease.To identify eligible original articles, we searched a series of computerized databases, including Medline via PubMed, EMBASE, Web of Science, and CNKI with a systematic searching strategy. The characteristics of each article and pooled odds ratios (ORs) with corresponding confidence intervals (CIs) were calculated and subgroup analysis was performed to analyze the source of heterogeneity.P < 0.05). The pooled data was calculated under the random-effects model as a significant moderate heterogeneity was found among the meta-analysis.A total of 2699 patients from seven articles were included in the meta-analysis. We identified a significant relationship between obstructive sleep apnea syndrome and gastroesophageal reflux disease, with a pooled OR of 1.75 (95% CI 1.18–2.59, The meta-analysis showed that there was a significant correlation between obstructive sleep apnea hypopnea syndrome and gastroesophageal reflux disease. Obstructive sleep apnea hypopnea syndrome (OSAHS) refers to apnea and hypopnea caused by partial or complete obstruction of upper airway collapse during sleep, usually accompanied by snoring, disturbed sleep architecture, decreased frequent oximetry values, daytime sleepiness, and inability to concentrate, that may further lead to multiple organ damages such as coronary heart disease, hypertension, type 2 diabetes mellitus, and dyslipidemia –3. It isIn the past decade, a large number of studies were conducted try to find the mechanism of OSAHS and GERD and establish a causal relation between them. Treatment with continuous positive airway pressure (CPAP) can improve the sleep quality in OSAHS patients and has been shown to improve regurgitation . A recenWe thus performed a meta-analysis to grade the strength of evidence and systematically explore whether OSAHS correlates with GERD in the literature (supporting information: PRISMA Checklist) .Studies reporting the obstructive sleep apnea hypopnea syndrome and gastroesophageal reflux were identified for inclusion. To identify eligible original articles, we searched a series of computerized databases, including Medline via PubMed, EMBASE, Web of Science, and CNKI using the following key terms: “gastroesophageal reflux,” “reflux,” “obstructive sleep apnea hypopnea syndrome,” “OSAHS,” “obstructive sleep apnea,” “OSA” separated by the Boolean operator AND or OR. Articles were searched in the computerized databases up to January 2018, without limits of language. Reference lists from the resulting publications were used to identify further relevant publications. We screened the titles and abstracts of the identified studies, and articles that could contain data regarding GRED and OSAHS were evaluated the full article. Two authors (ZH Wu and XP Yang) independently searched for papers and screened the reference lists of retrieved articles to further identify potentially relevant publications. Discrepancies were resolved by consensus.Original studies were carefully checked. There were no country restrictions. The inclusion criteria were (1) sleep disorders including OASHS; (2) studies in which all study group members exclude any possible predisposing factor (such as Barrett’s esophagus and asthma) that may be related to the development of their OSAHS; (3) studies that were restricted to humans, published in English and non-English, contained original data, and appeared in either abstract form or full text; (4) studies that clearly defined the study and control groups and the members, and observational data were available; and (5) studies that reported OSAHS patients with validated OSAHS (such as Epworth Sleepiness Score and Berlin questionnaire) questionnaire, polysomnography, and 24-h pH monitoring to document the acid reflux or anti-reflux treatment. The exclusion criteria were (1) case reports, non-English, abstracts, comments, review articles, duplicate publications, and editorials; (2) other treatment of obstructive sleep apnea (surgery or device); (3) studies that focus on laryngopharyngeal reflux (LPR) disease.Information was collected for each publication concerning the author’s name, publication year, study design, age, method of reflux evaluation or OSAHS evaluation, and study and control group criteria.P value of the chi-squared statistic when P was < 0.05. Higgins I2 test were used to assess the heterogeneity. An I2 value between 25 and 50% was considered as low heterogeneity, an I2 value between 50 and 75% as moderate heterogeneity, and an I2 value > 75% as high heterogeneity. An I2 value < 25% was considered homogeneous. When I2 value > 50%, the random-effects model was applied to combine effect size and when I2 value < 50% the fix-effects model was applied to combine effect size. We also sought to perform subgroup analysis to determine the sources of heterogeneity. The pooled odds ratios (ORs) of different articles and corresponding 95% confidence intervals (CIs) were used to estimate the relationship between OSAHS and GERD. The sensitivity analysis was repeated to assess the effects of individual study on pooled estimates by removing individual study.We used the PRISMA statement  to assesOur search strategy identified 1412 potentially relevant articles from electronic databases and 1 from reference lists and other sources. After excluding duplicates, 1298 records remained. Reading the titles and abstracts of these 1298 references led us to exclude 1218 articles that did not meet the inclusion criterion. After reading the full text of the remaining 80 articles as possibly reporting the relationship between OSAHS and GERD, 66 were excluded because the lack of sufficient data; 2 study were excluded because the outcome measure is inappropriate; 3 lacked basic information; 1 study was duplicate and we could not retrieve 1 study for full text. Ultimately, seven eligible articles were identified . The selThe meta-analysis consisted with a total sample size of 2699 included one case–control study and six cross-sectional studies. Two articles used 24-h pH monitoring to evaluate the relationship between OSAHS and GERD in cases and controls. Four articles used polysomnography to evaluate the apnea-hypopnea indices.P < 0.05). The pooled data was calculated under the random-effects model as a significant moderate heterogeneity was found among the studies.The forest plot result for association of OSAHS with GERD is shown in Fig. P < 0.05) under the random-effects model. The forest plot about the subgroup analysis is showed in Fig. The results was significant, with a corresponding value of 1.79  under the fixed-effects model. The forest plot about the subgroup analysis is showed in Fig. 4.The results was significant, with a corresponding value of 1.53  under the random-effects model. The forest plot about the subgroup analysis is showed in Fig. The results was significant, with a corresponding value of 2.73  under the fixed-effects model. The forest plot about the subgroup analysis is showed in Fig. 6.The results was significant, with a corresponding value of 1.44  under the fixed-effects model. The forest plot about the subgroup analysis is showed in Fig. 7.The results was significant in reflux questionnaire subgroup analysis, with a corresponding value of 1.53  under the fixed-effects model. The forest plot about the subgroup analysis is showed in Fig. The results was significant in 24-h pH monitoring subgroup analysis, with a corresponding value of 5.83 and when we removed Chen’s study, the results dramatically influenced the pooled results (I2 decreased from 54 to 3%) in the meta-analysis with the random-effects model. The outcome of sensitivity analysis showed that the pooled ORs ranged from 1.52 (95% CI 1.22–1.89) to 2.10 (95% CI 1.19–3.69). Moreover, in subgroup analyses, I2 = 67% for GERD versus non-GERD; I2 = 47% for OSAHS versus non-OSAHS; I2 = 0% for Berlin Questionnaire; I2 = 73% for polysomnography; I2 = 0% for reflux questionnaire; and I2 = 0% for 24-h pH monitoring. However, in GERD versus non-GERD subgroup, the results dramatically are influenced by removed Chen’s study (I2 decreased from 67 to 0%), and in OSAHS versus non-OSAHS subgroup, when we removed Basoglu’s study, the I2 decreased from 47 to 0%.The coupled forest plots show moderate heterogeneity . The pooled results were credible as the subgroups were analyzed. Although the value of I2 = 54% , it indicated that there existed moderate heterogeneity among the studies. But we performed the subgroup analysis to determine the sources of heterogeneity so the results of meta-analysis could represent the true relationship between OSAHS and GERD. Meanwhile, sensitivity analysis showed that after any individual study was omitted or the random-effects model was converted to a fixed-effects model, the overall results and conclusions still exist. Therefore, we have strong confidence to believe the result of our meta-analysis that strong association between OSAHS and GERD.This meta-analysis suggested that OSAHS was significantly associated with increased risk of GERD  in our meta-analysis under the random-effects model. Thus, we can conclude that GRED may play a very important role in the pathophysiology of OSAHS. OSAHS may be related to age, gender, BMI, alcohol, or Barrett’s esophagus (BE) in GRED patients. Ju et al. [Recent researches have reported that OSAHS and symptomatic in GERD are closely related. However, the exact causative relationship between them remains contradictory and this is the first meta-analysis to approve it. In Basoglu et al.  study, iu et al.  evaluateu et al. , 24. ThuP < 0.05) for polysomnography subgroup analysis with the random-effects model and the summary OR was 1.44  for Berlin Questionnaire subgroup analysis with the fixed-effects model. The subgroup analysis results indicated that GRED had a strong association with OSAHS. The standard diagnostic of GERD is 24-h dual-probe esophageal pH monitoring, and it can record episodes of not only laryngopharyngeal reflux but also gastroesophageal reflux. But in our meta-analysis, we only contain the data from gastroesophageal reflux. Many of patients (OSAHS in particular) maybe did not accept this application, as this is an invasive method. So few studies used this application assess the relationship between the OSAHS and GRED patients. We can only include two studies in our meta-analysis, and the summary OR was 5.83 for 24-h pH monitoring subgroup analysis under the fixed-effects model. These results supported that GERD participation in OSAHS pathogenesis.Usually, polysomnography is the standard diagnostic modality for OSAHS, but this procedure requires an overnight evaluation, so more studies have used questionnaires to evaluate patients’ sleep quality (such as the Berlin Questionnaire) . HoweverFrom this meta-analysis results, we may arrival at a conclusion that the GERD participation in OSAHS pathogenesis. On one hand, for patients with GERD, we can try anti-reflux treatment (such as proton pump inhibitors) and lifestyle modifications . On the other hand, we may also suggest that OSAHS patients be evaluated for GERD before undergoing surgical treatment or CPAP treatment. As an otorhinolaryngologist, if we suspected OSAHS patients with GERD signs and symptoms , fiberoThere are some potential limitations in the meta-analysis that should be squared up when interpreting the results of our study. Firstly, the sample size was relatively not enough and may affect the accuracy of our results and much large-scale studies should be performed to convince it. Secondly, although we try to explore the source of heterogeneity by subgroup analysis in our study, we could not successful explore heterogeneity from other aspects because of the insufficient clinical data and the limited studies. Thirdly, the results may also be biased by different measurement techniques to diagnosis OSAHS (polysomnography/Berlin Questionnaire) or GERD (reflux questionnaire/24-h pH monitoring). Additionally, as most of the included study population was adult, the results may be different in children population. So our results should be interpreted with caution and need further researches. Despite there are limitations, our analysis shows a strong and clear association between OSAHS and GERD.This meta-analysis provided direct evidence of the GERD participation in OSAHS pathogenesis and suggested that in treatment of OSAHS, the GERD disease should not be neglected in clinical practice.


Background and Objectives: The ACE gene encodes the angiotensin-converting enzyme (ACE), a component of the renin-angiotensin system. Increased ACE activity may cause abnormal regulation of placental circulation and angiogenesis, resulting in adverse pregnancy outcomes. Previous studies have reported that the insertion/deletion (I/D) polymorphism of the ACE gene is associated with the development of preterm birth (PTB). However, results of the association between ACE gene I/D and PTB are inconsistent in various populations. Therefore, we performed a case-control study and a meta-analysis to evaluate the association between ACE I/D polymorphism and PTB. Materials and Methods: We analyzed a total of 254 subjects (111 patients with PTB and 143 women at ≥38 weeks gestation) for the case-control study. For the meta-analysis, we searched Google Scholar, PubMed, and NCBI databases with the terms “ACE,” “angiotensin-converting enzyme,” “preterm birth,” “preterm delivery,” and their combinations. Results: Our results of the case-control study indicated that ACE I/D polymorphism is significantly associated with PTBs in the overdominant genetic model (odds ratio (OR) 0.57, 95% confidence interval (CI) 0.347–0.949, p = 0.029) and that the ID genotype of ACE I/D polymorphism has a protective effect for PTB . Similarly, the meta-analysis showed that the OR for the ACE gene ID genotype was 0.66 . Conclusion: The ACE gene ID genotype has a significant association with PTB and is a protective factor for PTB. A larger sample set and functional studies are required to further elucidate of our findings. However, the pathogenesis of PTB is not well understood [Preterm birth (PTB) is a pregnancy-related disorder defined as a birth before 37 weeks of gestation and is known to increase the risk of infant mortality . AccordiVarious factors, such as urogenital infection, abnormal placentation, gestational bleeding, and uterine and cervical anomalies, increase the risk of PTB . SeveralACE gene is located on chromosome 17q23.3 and has 26 exons and 25 introns [ACE is generally associated with the level of ACE concentration [The human  introns . A 287 bntration . The DD ntration . Increasntration .ACE gene I/D polymorphism and gestational disorders. Valdez et al. (2007) found that the D allele of ACE gene I/D polymorphism was significantly associated with premature rupture of membranes in the Mexican population [ACE gene I/D polymorphism and uteroplacental dysfunction in the Middle Eastern Asian population [ACE gene I/D polymorphism was associated with the development of PTB in the Caucasian population [ACE gene I/D polymorphism in the Turkish population [A number of studies have analyzed the genetic association between pulation . Häupl epulation . Uma et pulation . Howeverz et al. 07 found pulation .ACE gene I/D polymorphism. A meta-analysis offers an opportunity to quantitatively review overall association data from individual results of the same genetic polymorphism [ACE gene I/D polymorphism and PTB in Korean women and to corroborate the results using a meta-analysis.Therefore, we conducted a case-control study and a meta-analysis to investigate the genetic association between preterm birth and morphism . A meta-morphism . This stIn this study, a total of 254 women were recruited from the Obstetrics and Gynecology Department at Dankook University Hospital in Korea. One hundred and eleven PTB patients were selected, where the gestational ages were 24–37 weeks . One hunACE gene I/D polymorphism were detected using PCR. Primer sets were designed according to the previous study reported by Gürdöl et al. [2, 10 × PCR buffer, and 1.0 U NV DNA polymerase . The PCR amplification was conducted with a C1000 Touch thermal cycler  under the following conditions: 95 °C for 5 min, followed by 40 cycles at 94 °C for 1 min, 64 °C for 1 min, and 72 °C for 1 min, then a final extension at 72 °C for 10 min. Fragments with insertion  and without insertion  were detected as 490 and 190 bp, respectively, in 2% agarose gel .DNA was extracted from leukocytes or buccal cells using the GeneAll Exgene Clinic SV mini kit . Genotypes of l et al. : forwardl et al. . The PCRACE gene I/D polymorphism and PTB. The PubMed and Google Scholar databases were used as search sources, with the keywords “ACE,” “angiotensin-converting enzyme,” “preterm birth,” “preterm delivery,” “preterm labor,” “premature delivery,” and their combinations. We used the following criteria for selection: English language, human population-based case-control studies, studies for association of ACE gene I/D polymorphism and preterm birth, studies consisting of detailed genotype data with proof of Hardy–Weinberg equilibrium (HWE), and studies for only maternal subjects, excluding meta-analysis. Finally, we selected three studies published from 2007 to 2008.We performed a meta-analysis of the published literature and present study to analyze the association between ACE gene I/D genotypes.We extracted the following data from each study: last name of the first author, publication year, subjects’ ethnicity, sample size, number of cases, and control for t-test was performed to compare the following data characteristics: age, height, weight, pre-gestation weight, systolic blood pressure, diastolic blood pressure, birth weight, and gestational age at delivery using SPSS software . Chi-squared tests were used to assess HWE. In addition, odds ratios (OR) with 95% confidence intervals (CIs) were calculated in a 2 × 2 table using the genotype and allele frequencies of cases and controls using web-based statistics tools (SISA (http://www.quantitativeskills.com/sisa/) and SNPStats (http://bioinfo.iconcologia.net/SNPStats)). For the meta-analysis, the chi-squared test was performed to prove whether the genotype frequencies in the control group of each reference were assessed for HWE. ORs of each study and pooled data with 95% CIs were calculated to analyze the association between ACE gene ID genotype and PTB in the overdominant genetic model with fixed-effect and random-effect models [I2, i.e., the percentage of the variability [I2 value was over 50%, indicating significant heterogeneity between studies, we selected a random-effect model. If under 50%, we selected a fixed-effect model for analysis to analyze the general effect of the ACE gene ID genotype. To investigate evidence for publication bias, we used the test for funnel plot asymmetry described by Egger et al. [p value of <0.05 was considered statistically significant in all statistical analyses.An independent t models . Becauseiability . If the r et al. . In the r et al. . All metp > 0.05). However, a significant difference was found in birth weight and gestational age between PTB patients and controls (p < 0.05) , and DBP (diastolic blood pressure), no meaningful differences were observed between the PTB patients and controls ( < 0.05) .ACE gene I/D polymorphism for the 111 patients with PTB and 143 controls are presented in ACE gene I/D polymorphism did not deviate from the Hardy–Weinberg equilibrium (p = 0.043). In the overdominant genetic model, we also found significant association between the ID genotype of ACE gene I/D polymorphism and PTB .Genotyping data of the ilibrium . The ID ACE I/D polymorphism, six did not follow the Hardy–Weinberg equilibrium (HWE) in control groups (p < 0.05), one used only fetal subjects, and one duplicated the subjects. Finally, we selected four studies with 467 patients and 365 controls, including the present case-control study [A total of 684 papers were initially searched from the Google Scholar and PubMed databases with the related keywords . Thirty-ol study ,19,30. TACE gene I/D polymorphism is associated with PTB. Therefore, we conducted a meta-analysis for the overdominant genetic model to corroborate this result. The OR for the fixed-effect model was 0.66 , indicating that the ACE gene ID genotype is protective to PTB and the data indicated no significant heterogeneity between respective studies  (Our case-control study showed that the ID genotype of  = 0.83) . The pub = 0.83) . Each ofACE gene I/D polymorphism on PTB. The results of a case-control study showed that the ACE gene ID genotype is protectively associated with the development of PTB in Korean women    . We also= 0.029) .ACE gene I/D polymorphism has a protective effect  (To provide a more precise estimation of the association between < 0.008) .ACE gene with PTB should be determined, but we analyzed only the genotype data from mothers. Finally, PTB is a complex disease caused by the interaction between many factors [ACE gene I/D polymorphism and PTB.The present study should be interpreted with the following limitations: first, the sample size in this case-control study was not large enough to represent the whole of the Korean population and detect reliably significant associations . A large factors ,however,ACE gene I/D polymorphism and PTB has not been extensively researched [Despite these limitations, our study has many advantages. The association of searched ,21, so oACE gene I/D polymorphism has a significantly protective effect on the pathogenesis of PTB. Similarly, the subgroup analysis also indicated that the polymorphism was related to a decrease in the incidence of PTBs. However, as PTB is multifactorial, our findings require verification with larger sample sizes and functional studies.Our results suggest that the ID genotype of


The present study aims to investigate the incidence and risk factors associated with postoperative delirium in patients undergoing spine surgery.  PubMed, EMBASE, Cochrane Library, and Science Citation Index were searched up to August 2019 for studies examining postoperative delirium following spine surgery. Incidence and risk factors associated with delirium were extracted. Odds ratios (OR) and 95% confidence intervals (CI) were calculated for outcomes. The Newcastle–Ottawa Scale (NOS) was used for the study quality evaluation.  The final analysis includes a total of 40 studies. The pooled analysis reveals that incidence of delirium is 8%, and there are significant differences for developing delirium in age , age more than 65 , age more than 70 , and age more than 80  years, male , a history of alcohol abuse , anxiety , congestive heart failure , depression , hypertension , kidney disease , neurological disorder , opioid use , psychoses , pulmonary disease , higher mini-mental state examination , preoperative pain , and postoperative urinary tract infection .  A comprehensive understanding of incidence and risk factors of delirium can improve prevention, diagnosis, and management. Risk of postoperative delirium can be reduced based upon identifiable risk factors. Postoperative delirium is a common complication after surgery in the elderly and causes difficulty in postoperative care , 2. It iThe systematic review and meta-analysis were done according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guideline and AMSTAR  Guidelines , 6. PubMStudies included in this systematic review and meta-analysis met the following criteria: (1) original articles on patients who underwent spine surgery, (2) observational, case series or cohort study design, (3) at least incidence reported or one risk factor identified as being associated with delirium, and (4) full text available. If the inclusion criteria were not met, the study was excluded. If the same study was published in different years or various journals, then the most frequently cited study was included for this meta-analysis. The potentially qualified studies were selected independently by 2 authors according to the inclusion and exclusion criteria. Any discrepancy was resolved by discussion to reach a consensus.Data were extracted by two independent authors. By discussion or by involving a third author, disagreements were addressed. The general features cover the first author, publication year, country, study type, sample size, patient characteristics, patients who underwent surgery, delirium diagnosis tool, incidence duration of delirium, and significant factors.Two authors independently evaluated the quality of the studies, and the level of agreement between them was recorded. Any disagreements between the 2 authors were resolved by discussion with a third author. Newcastle–Ottawa Scale (NOS) was utilized to assess the quality of each study  since noI2 statistic, and an I2 of >50% was considered to indicate substantial heterogeneity. The random-effects model or the fixed-effect model was used depending on the heterogeneity of studies included. A random-effects model was used for heterogeneous data. Otherwise, a fixed-effect model was used. Begg's and Egger's test were used to estimate publication bias, when 10 or more studies are presented. For any variable presenting with large heterogeneity, sensitive analysis or subgroup analysis was used to investigate the potential origin of heterogeneity.The meta-analysis of comparable data was performed using Stata/SE version 15.0 software. All adjusted odds ratio (OR) with 95% confidence interval (CI) were collected and pooled to evaluate the relationships between various risk factors and postoperative delirium in patients undergoing spine surgery. In addition, crude ORs with 95% CIs were calculated based on the frequency reported in the original literature. Inconsistency was quantified with There were 1360 relevant studies included according to the search strategy. After the titles and abstracts were reviewed, 1256 of them were removed. A full-text review was evaluated in the 104 records maintained, and 64 of them were excluded because they did not meet the inclusion criteria. Finally, 40 studies representing 712820 patients were included in the present meta-analysis .The characteristics of the included studies are summarized in I-squared > 50%, P < 0.001). Interestingly, the heterogeneity remained high with each of the subgroups of study type, countries, or operated levels (P < 0.001)  .The ORs and 95% CIs of the risk factors are displayed in After synthesis of 7 studies, it revealed that patients who developed delirium were significantly older . Meanwhile, age older than 65 , 70 , and 80  years were significantly associated with the risk of developing delirium. Another demographic factor male was considered to be associated with less delirium risk in the pooled analysis .A history of alcohol abuse , anxiety , congestive heart failure , depression , hypertension , kidney disease , neurological disorder , opioid use , psychoses , and pulmonary disease  were more likely to develop delirium than controls. Assessment of mental state, as measured by mini-mental state examination (MMSE), demonstrated a significantly lower risk to develop delirium in patients with higher scores . In addition, preoperative pain and postoperative urinary tract infection (UTI) were related to the development of delirium .Delirium is thought to be a less transient disorder than previously believed in several studies , 11. In This systematic review and meta-analysis were performed to pool and identify the incidence and risk factors of postoperative delirium after spine surgery. The pooled incidence of delirium in this meta-analysis is 8%. However, the present study showed wide variation and heterogeneity in incidence of delirium. A previous meta-analysis of 6 studies reported incidence of delirium after spine surgery varies from 0.84% to 21.3% . InteresOne of the most important risk factors was older age, especially in patients over 65. This may be attributed to the fact that elderly patients are more likely influenced by age-related physical and psychical changes. Aging is also associated with a higher incidence of comorbidity such as hypertension, diabetes mellitus, and pulmonary disease , 30. TheI2 0%)  . Only onI2 0%) . InteresRegarding the comorbidity of hypertension, the meta-analysis of 13 studies , 45, 46 I2 0%). Meanwhile, parkinsonism was also not found to be related to postoperative delirium . However, there is still controversy in the role of parkinsonism for postoperative delirium. Kim et al. [I2 88%). After subgroup analysis, there was a high heterogeneity between retrospective studies . Through subgroup analysis, we found that geographical factors may contribute to heterogeneity . This meThe finding that preoperative pain and opioid use is associated with increased probability of delirium has been previously reported in patients with or without hip fracture or patients with cancer , 50. In In our study, intraoperative factors do not appear to influence the prevalence of delirium based on normal clinical practice such as blood loss, blood transfusion, cervical surgery, dural tear, operated levels, and operation time. Notably, for intraoperative blood loss, there was high heterogeneity among studies . After sThere are some limitations in our study. First, no randomized controlled trials were included despite our exhausted search from literatures, which may influence the quality of the result. Second, although subgroup analyses were used, the pooled result of incidence was still reported with high heterogeneity, which should be explained with caution.In summary, the study reveals that pooled incidence of delirium is 8% and age, gender, history of alcohol abuse, anxiety, congestive heart failure, depression, hypertension, kidney disease, neurological disorder, opioid use, psychoses, pulmonary disease, MMSE, preoperative pain, and postoperative UTI were significant factors for delirium after spine surgery. A comprehensive understanding of incidence and risk factors of delirium can improve prevention, diagnosis, and management.