Anemia in pregnancy is a serious threat to maternal and child health and is a major public health problem. However, the risk factors associated with its incidence are unclear and controversial.PubMed, Ovid Embase, Web of Science, and Cochrane databases were systematically searched . The screening of search results, extraction of relevant data, and evaluation of study quality were performed independently by two reviewers.A total of 51 studies of high quality (NOS score ≥ 7) were included, including 42 cross-sectional studies, six case-control studies, and three cohort studies. Meta-analysis showed that infected parasite, history of malarial attack, tea/coffee after meals, meal frequency ≤ 2 times per day, frequency of eating meat ≤ 1 time per week, frequency of eating vegetables ≤ 3 times per week, multiple pregnancies, multiparous, low household income, no antenatal care, rural residence, diet diversity score ≤ 3, have more than 3 children, history of menorrhagia, underweight, family size ≥ 5, middle upper arm circumference < 23, second trimester, third trimester, birth interval ≤ 2 year were all risk factors for anemia in pregnancy.Prevention of anemia in pregnancy is essential to promote maternal and child health. Sufficient attention should be paid to the above risk factors from the social level and pregnant women's own aspects to reduce the occurrence of anemia in pregnancy.https://www.crd.york.ac.uk/prospero/, identifier: CRD42022344937. As a global public health problem, anemia in pregnancy has been shown to be an independent risk factor for adverse maternal and infant outcomes such as blood transfusion, postpartum hemorrhage, cesarean section, hysterectomy, preterm birth, and infectious diseases . It direIt is critical to explore the risk factors that may cause anemia in pregnancy and take preventive strategies as soon as possible. However, the risk factors for anemia in pregnancy are controversial. For example, the findings of Kedir et al. suggest that parasite infection is not a risk factor for anemia in pregnancy . HoweverThe current field lacks definitive evidence on the risk factors for anemia in pregnancy. Therefore, as the first study to systematically summarize the risk factors of anemia in pregnancy, the results of this study can provide a reference for the prevention and treatment of anemia in pregnancy in the future.https://www.crd.york.ac.uk/prospero/).This systematic review and meta-analysis followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines . The revPregnant women.Report at least one exposure factor associated with anemia in pregnancy.Studies where adjusted odds ratio (AOR) for exposure factors were available or calculated.Anemia in pregnancy occurs. The diagnostic criteria are hemoglobin ≤ 11 g/dL.Cross-sectional studies, case-control studies, and cohort studies.Animal studies and cell experiments were excluded. Reviews, case reports, opinion articles, conference abstracts, and non-published data were also excluded.Candidate studies were identified through searches of the PubMed, Web of Science, Cochrane, and Ovid Embase databases from inception until June 27, 2022. Also, the reference lists of the included studies were searched. The retrieval approach of the combination of free words and subject words was adopted. The following terms were combined to generate search keywords: [gestational anemia OR anemia in pregnancy OR (pregnancy OR pregnant OR gestation) AND anemia] AND (hazard OR risk factors OR risk factor OR related factors OR factors OR influence factors OR influencing factors). Further details of the search strategy are shown in Literature screening and data extraction were performed by 2 trained researchers according to the inclusion and exclusion criteria as indicated previously. Extracted content includes: (1) Basic information of included studies: authors, year, country, type of study, sample size, age, method of obtaining information, diagnostic criteria for anemia, and data analysis methods. (2) Exposure factors: Risk factors related to dietary habits, self-condition, and disease history of pregnant women. (3) Key elements of risk of bias assessment.Based on the Newcastle-Ottawa Scale (NOS), two qualified researchers independently evaluated the inherent risk of bias of included studies from three aspects, including the selection of participants, confounding variables, and measurement of exposure . The eva2 test was used to evaluate the heterogeneity of the included studies (the test level was α = 0.05), and the size of the heterogeneity was judged according to the I2 value. When P > 0.05 and I2 ≤ 50%, it indicated that the heterogeneity of the results of each study was not statistically significant, and a fixed-effects model was used for meta-analysis; otherwise, after further analysis of the source of heterogeneity, a random-effects model was used.Statistical analysis was performed using STATA 16 software. Results were calculated using adjusted odds ratio (AOR) and 95% confidence interval (95% CI). The χA total of 4,638 relevant records were obtained from the initial inspection, which were excluded from repeated studies, non-risk factor studies , pregnant women without anemia  and study types that are not consistent , 51 studies were finally included. The article screening process is shown in A total of 51 studies were included , 18–60, The NOS scores of the 51 included studies were all ≥ 7 points, of which 38 studies had a NOS score of 8 points, and 13 studies had 7 points, indicating that the included studies had high research quality. See I2 = 0, P > 0.05, indicating that there is little possibility of heterogeneity among the studies, a fixed-effect model was used for combined analysis. Meta-analysis showed that parasitic infection  and history of malarial attack  were risk factors for anemia in pregnancy, while HIV status  and abortion history  were not associated with anemia in pregnancy  As the first study in the current field to systematically summarize the risk factors of anemia in pregnancy, this study may serve as the best evidence for the prevention of anemia in pregnancy in the future. (2) This study was based on AOR rather than OR analysis, avoiding the interaction between multiple exposure factors, and the results were more in line with the actual situation. (3) According to the NOS scoring results, we found that the quality of evidence of the 51 included studies was high, which ensured the reliability of the meta-analysis results.Limitations: (1) The included studies were all from developing countries, especially Ethiopia, therefore, our findings are only applicable to some countries, not all countries. (2) Include only English literature, which may lead to language bias. (3) The gray literature and conference abstracts were not searched, which may lead to publication bias.The high incidence and serious harm of anemia in pregnancy make it urgent to systematically summarize its risk factors. Evidence from 51 high-quality studies showing infected parasite, history of malarial attack, tea/coffee after meals, meal frequency ≤ 2 times per day, frequency of eating meat ≤ 1 time per week, frequency of eating vegetables ≤ 3 times per week, multiple pregnancies, multiparous, low household income, no antenatal care, rural residence, diet diversity score ≤ 3, have more than 3 children, history of menorrhagia, underweight, family size ≥ 5, middle upper arm circumference < 23, second trimester, third trimester, birth interval ≤ 2 year, these 20 exposure factors were all risk factors for Anemia in Pregnancy. Therefore, health institutions and pregnant women themselves should focus on the above risk factors for better prevention and early detection of anemia in pregnancy.The original contributions presented in the study are included in the article/YSu proposed ideas and designed protocol. JZ and QL were responsible for data analysis and writing of the paper. YSo, LF, and LH were responsible for literature screening, data extraction, and quality evaluation. All authors contributed to the article and approved the submitted version.We acknowledge the financial support from Kunming Health Science and Technology Personnel Training Project (No. 2022-SW-93).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.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


Pancreatic ductal adenocarcinoma (PDAC) has a poor prognosis, and this is attributed to it being diagnosed at an advanced stage. Understanding the pathways involved in initial development may improve early detection strategies. This systematic review assessed the association between circulating protein and metabolite biomarkers and PDAC development.A literature search until August 2020 in MEDLINE, EMBASE, and Web of Science was performed. Studies were included if they assessed circulating blood, urine, or salivary biomarkers and their association with PDAC risk. Quality was assessed using the Newcastle-Ottawa scale for cohort studies. Random-effects meta-analyses were used to calculate pooled relative risk.n = 4 studies; pooled relative risk (RR): 1.61; 95% CI: 1.16–2.22]. Additionally, an inverse association was seen with pyridoxal 5′-phosphate (PLP) levels . Meta-analyses showed no association between levels of C-peptide, members of the insulin growth factor signaling pathway, C-reactive protein, adiponectin, 25-hydroxyvitamin D, and folate/homocysteine and PDAC risk. Four individual studies also reported a suggestive positive association of branched-chain amino acids with PDAC risk, but due to differences in measures reported, a meta-analysis could not be performed.A total of 65 studies were included. Higher levels of glucose were found to be positively associated with risk of developing PDAC [Our pooled analysis demonstrates that higher serum glucose levels and lower levels of PLP are associated with risk of PDAC.Glucose and PLP levels are associated with PDAC risk. More prospective studies are required to identify biomarkers for early detection. Pancreatic cancer is considered to be one of the most lethal cancers with a very poor prognosis, and only 5% of patients survive for 10 or more years after diagnosis . This isThe incidence of PDAC in the general population is low and accounts for only 3% of all new cancer cases diagnosed in the UK . This maA majority of the studies on circulating biomarkers associated with PDAC biology are case–control studies that have assessed the biomarker very close to diagnosis and therefore do not provide a lot of information on the initial development of the cancer . This reTo assess the association between circulating biomarkers and risk of developing PDAC, Medline (1974–), Embase (1974–), and Web of Science (1970–) were searched systematically for eligible studies in humans using predefined search terms from date of inception to May 10, 2019. Medical Subject Headings and keywords for cancer, biomarkers, sample , and early detection/risk were used.An updated search was performed in all three databases on August 20, 2020, to identify any new articles published before beginning the final analysis. The detailed search strategy for each database is included in Supplementary Table S1, and this protocol was registered in the PROSPERO database . Any discrepancies were resolved by discussions with a third reviewer and a consensus was reached.Titles and abstracts of the studies identified by the search were screened for eligibility by two independent reviewers  based circulating biomarkers and risk of subsequent development of PDAC. Included studies were required to have a follow-up period of at least six months after biomarker assessment and report on both the measure of association, in terms of odds ratios/hazard ratios and their corresponding 95% confidence intervals, or have enough data for these to be calculated. Case–control/retrospective studies, studies where the biomarker was measured at or close to the time of diagnosis  and those on biomarkers of cancer mortality were excluded. Participants/population included anyone with a diagnosis of pancreatic cancer or individuals who had no history of cancer at the time of biomarker assessment.Because the main focus of the review was to look at protein/metabolite markers, studies assessing blood-based genetic markers, miRNA, and infectious agents were not included in the final analysis, and only abstracts were screened for the biomarker studied.Full texts of the articles were assessed for eligibility, and data extraction was performed by one reviewer (SK). Extracted data from the individual studies included author names, date of publication, study characteristics, participant details, biomarker assessed, sample type, and outcomes of interest. The data extraction was checked by a second reviewer  to ensure accuracy. Study quality was assessed by using the Newcastle-Ottawa scale for cohort studies .2 test was used to investigate heterogeneity and I2 statistic was calculated to report on variation between the study estimates. Heterogeneity was considered high if I2 statistic was above 75% .I2 = 76%; I2 = 0%) or Jee and colleagues (ref. I2 = 0%) lowered the heterogeneity observed and the positive association remained. Two studies reported consistent positive association between circulating levels of insulin : 1.61; 95% CI: 1.16–2.22,  insulin  and HbA1 insulin  with PDAI2 = 36%; I2 = 0%; I2 = 26%; We also identified four studies that looked at circulating levels of various components of the insulin growth factor (IGF) axis and their association with PDAC risk. The meta-analyses showed no significant association between levels of IGF-1 .We identified 17 studies that assessed the association between different nutrition-related biomarkers and PDAC risk, and these are listed in Supplementary Table S4. A meta-analysis of four studies found a significant inverse association between levels of circulating pyridoxal 5′-phosphate (PLP), which is the active form of vitamin B6 and risk of PDAC development -D) and risk of developing PDAC ; We identified 17 studies that looked at the association between metabolism-related biomarkers and PDAC risk (Supplementary Table S5). A meta-analysis of five studies found a trend toward an inverse association between levels of total serum cholesterol and risk of PDAC [pooled relative risk (RR): 0.89 95% Confidence Interval (CI): 0.79–1.00, I2 = 61%; I2 = 75%; We also identified four studies that assessed the role of metabolites involved in the one-carbon metabolite pathway in the early development of PDAC. A meta-analysis of three studies found no association between the levels of folate . A meta-analysis of three studies looking at circulating adiponectin levels found no association with PDAC risk  and PDAC risk. A meta-analysis found no association between the levels of this marker and pancreatic cancer risk . Additionally, an inverse association was found between levels of cholesterol (n = 5 studies) and PLP (n = 4 studies).The aim of this systematic review was to assess the association of various circulating biomarkers with PDAC risk, in order to understand their role in the early development of this cancer. We identified 65 eligible articles, and meta-analysis showed a positive association between glucose levels and PDAC risk  and PDAC risk in a meta-analysis of four studies . All fouThe main strength of this review is that, to the best of our knowledge, it is the first comprehensive systematic review and meta-analysis conducted on the association of circulating biomarkers as a whole and PDAC risk. The quality and risk of bias was assessed using the established tool NOS  and seveIn summary, our findings strengthened the evidence on the role of increasing glucose levels with PDAC development and also discovered an inverse association with PLP levels. We also identified a possible role of BCAAs and trend toward the role of low levels of cholesterol in the early development of PDAC. However, further research is required in order to draw definitive conclusions on the nature of these relationships and deepen our understanding of the role of these pathways in governing PDAC biology.S. Kumar reports grants and personal fees from European Union Horizon 2020 Marie Sklodowska-Curie Actions COFUND Doctoral training program during the conduct of the study. A.J. McGuigan reports grants from Royal College of Surgeons and Ulster Society of Gastroenterology during the conduct of the study. R.C. Turkington reports personal fees from Almac Diagnostics outside the submitted work. No disclosures were reported by the other authors.Supplementary DataClick here for additional data file.Supplementary DataClick here for additional data file.Supplementary DataClick here for additional data file.Supplementary DataClick here for additional data file.Supplementary DataClick here for additional data file.Supplementary DataClick here for additional data file.Supplementary DataClick here for additional data file.Supplementary DataClick here for additional data file.


Background and objectivePharyngocutaneous fistula is the most common complication after total laryngectomy. The aim of this study was to examine the incidence of fistula and the association between fistula and the shape of mucosal closure .MethodA search of English language databases from 1979 to 2021 was undertaken for studies of total laryngectomy that commented on pharyngeal closure techniques and fistulas. Pooled estimates for fistula incidence and odds ratios were calculated.ResultsA total of 24 retrospective studies were included. The pooled fistula rates in primary total laryngectomy were 19.9% with T-shaped closure, 16.1% with vertical closure, and 16.4% with horizontal closure. In salvage total laryngectomy, the pooled fistula rates were 35.1%, 36.1%, and 17.9% with T-shaped, vertical, and horizontal closure, respectively. In the analysis of association, the risk of fistula formation in the T-shaped closure was not significantly different compared to that in the vertical closure, . The horizontal closure, when compared to vertical closure, was significantly associated with lower risk of fistula formation , but had nonsignificant lower risk of fistula formation when compared to the T-shaped closure .ConclusionHorizontal closure seems to be the best closure shape for primary repair after total laryngectomy. However, analysis bias may have occurred because of the lack of well-controlled studies. Total laryngectomy is a commonly used curative procedure for the treatment of advanced and recurrent tumors, and other noncancerous conditions of the larynx. After complete removal of the larynx, the resulting defect of the pharynx is repaired, creating the so-called neopharynx. Ideally, the pharyngeal reconstruction must be watertight to avoid leakage, sufficiently large to allow food passage, and capable of accommodating voice rehabilitation. Complications related to neopharyngeal reconstruction include salivary fistula, diverticulum, and stenosis.Pharyngocutaneous fistula is the most common complication related to neopharyngeal reconstruction . PreviouWhen possible, primary closure is the first choice for neopharyngeal reconstruction because it has better function and is less complicated . The mucthis study aimed to pool the available data on the fistula incidence of each closure shape and to determine the association between these closures and risk of fistula formation.MethodsSearch StrategyA search of English language databases from 1979 to 2021 was undertaken. The studies were retrieved from The Cochrane Library, PubMed, and Google Scholar databases using the free text terms, “total laryngectomy,” “fistula,” and “closure.” Hand searches of the reference lists of the included articles were also performed.Selection CriteriaTwo authors independently assessed the titles and abstracts for eligibility, and full-text articles were reviewed for closer examination. Studies were included if the following criteria were fulfilled: (1) the study reported on fistula incidence after total laryngectomy with primary closure reconstruction, (2) types of closure and number of cases were mentioned in detail. (3) For pooled analysis of incidence, the fistula outcomes between primary or salvage surgery were clearly specified. (4) For association analysis, the study contained two or more comparative groups with different closure shapes. Exclusion criteria were as follows: (1) usage of primary closure with flap reinforcement, (2) usage of staple closure or salivary bypass tube, and (3) case reports and case series.Quality AssessmentFor pooled analysis of incidence, studies were graded using six criteria based on the STROBE statement . The criStatistical AnalysisMeta-analysis was performed using Review Manager software (version 5.2). The random effects and fixed effects models were used, but the random effects model was preferred because of the diversity of surgical techniques and patients among the included studies. Studies with low quality or high risk of bias were not included in the pooled analysis. The pooled incidence rates of fistulas after primary and salvage total laryngectomy were calculated separately. Differences of fistula outcome between closure shapes were considered significant if a 95% confidence interval did not include the null value.ResultsFigure Tables From the included studies, 33-89% of the fistula occurred after primary total laryngectomy was resolved with conservative treatments. In salvage total laryngectomy, the success rates of fistula closure with conservative treatments were 0-29%.Table Horizontal closure, when compared to vertical closure, was significantly associated with lower risk of fistula formation but had a nonsignificantly lower risk of fistula formation when compared to T-shaped closure. The low level of heterogeneity indicated that the majority of studies showed favorable outcomes toward horizontal closure.DiscussionPharyngocutaneous fistula is the most common complication encountered in the postoperative period following total laryngectomy. Surgical techniques during neopharyngeal reconstruction are crucial in preventing post-operative fistulas.Several issues regarding surgical techniques for pharyngeal reconstruction have been studied. For example, recent studies have shown that the use of continuous sutures is associated with lower fistula rates than interrupted sutures in mucosal repair, and nonclosure of the pharyngeal constrictor muscle is not associated with a higher rate of fistula formation ,30. HoweThis study shows that vertical closure yields similar fistula incidence rates and outcomes when compared with T-shaped closure. Both closure shapes have their advantages and disadvantages. The trifurcation in the T-shaped closure might increase the risk of fistula formation. However, the T-shaped closure causes less tension than the vertical closure in some defect shapes, which decreases the risk of fistula formation . The surSwallowing is another aspect to be considered. Previous studies have reported that the swallowing function of horizontal and T-shaped closures was superior to that of vertical closure ,31. The Horizontal closure seems to be the ideal closure shape for primary closure of the neopharyngeal reconstruction. Horizontal closure produces a relaxed and wide neopharynx, similar to T-shaped closure, but avoids trifurcation, similar to vertical closure. The results of this study also support the superior outcome of horizontal closure in fistula prevention. However, horizontal closure may have limited use because it is not suitable for vertically extended pharyngeal defects.The following limitations should be considered in the analysis. First, none of the included studies were controlled trials. Other known risk factors for fistula formation, such as age, smoking, T-stage, tumor site, and nutritional status, might confound analysis outcomes.Salvage surgery after concurrent chemoradiotherapy had higher complication and fistula rates than surgery after radiotherapy alone . In the Horizontal and vertical closures were usually performed in smaller pharyngeal defects. Therefore, bias related to the T-stage and tumor site might have occurred. Omitting nonEnglish-language studies might result in biased samples. Finally, although horizontal closure showed a significantly lower risk of fistula formation than vertical closure, only two studies were available for analysis. Thus, the results might be limited.The fistula outcomes between T-shaped and vertical closures were comparable. Horizontal closure had notably lower fistula incidence in salvage total laryngectomy and was associated with a lower risk of fistula formation when compared with T-shaped and vertical closures. These findings suggest that horizontal closure is the best closure shape for primary repair after total laryngectomy. However, bias in analysis may have occurred because of the lack of well-controlled studies.


Objective: The relationship between gene polymorphisms in long non-coding RNAs (LncRNAs) and cervical cancer susceptibility has been thoroughly analyzed; however, the conclusions are inconsistent. Therefore, this systematic review and meta-analysis aimed to accurately assess the relationship between them.Method: Eligible literatures were retrieved from PubMed, Medline, China National Knowledge Infrastructure (CNKI), and WanFang databases before 1 April 2022. The odds ratios with the corresponding 95% confidence intervals were used to evaluate the strength of these relationships. Sensitivity analysis for publication bias was conducted to assess the stability and reliability of included literatures.Results: A total of 59 SNPs in 11 LncRNAs were summarized for a systematic review in this study, and then, a meta-analysis of rs920778 and rs4759314 polymorphisms in HOTAIR and rs217727 polymorphisms in H19 was conducted. The results demonstrated that rs920778 and rs4759314 polymorphisms were significantly correlated with cervical cancer susceptibility. Further subgroup analysis of rs920778 polymorphism showed that both small sample size and large sample size subgroups were associated with cervical cancer susceptibility. However, no association was found between rs217727 polymorphism and cervical cancer risk in all five genetic models.Conclusion: In conclusion, the rs4759314, rs920778, and rs217717 polymorphisms of HOTAIR and H19 may be associated with cervical cancer. However, the results should be interpreted with caution due to the limited sample and heterogeneity in this study. Large-scale and well-designed studies need to be practiced to validate our results. Cervical cancer is the most common malignancy affecting the female reproductive system, and its incidence is increasing every year. It is the fourth most commonly diagnosed cancer worldwide and the leading cause of cancer deaths. According to statistics, there were approximately 570,000 new cases and 310,000 deaths worldwide in 2018 . In the Etiological studies have shown that cervical cancer is closely associated with human papillomavirus (HPV) infection, but most cases of HPV infection are self-limiting, and only a small proportion of HPV-infected individuals eventually develop cervical cancer . DiffereRecently, long non-coding RNAs (LncRNAs) have been identified as a group of molecules closely associated with cancer heterogeneity. LncRNAs are more than 200 nucleotides in length, include polyA tails, cap structures, and promoter structures, and do not have an open reading frame (ORF) . TherefoSingle nucleotide polymorphisms (SNPs) are among the most common genetic variants in the population, with approximately 10 million SNPs in the human genome . SNPs inIn this study, we conducted a systematic review and meta-analysis of SNPs in LncRNAs and cervical cancer susceptibility, and explored the pathogenesis of cervical cancer. SNPs in LncRNAs can be used as molecular markers to screen high-risk populations or susceptible individuals, thus providing a scientific basis for early diagnosis and treatment of cervical cancer in the future.A systematic search of PubMed, Medline, CNKI, and Wanfang databases was conducted. Literature with a publication date before 1 April 2022 was searched. The search keywords were “cervical or CIN,” “long non-coding RNA or LncRNA,” and “polymorphisms or polymorphism”.The following studies were included in this meta-analysis: 1) case-control studies; 2) studies in which patients in the case group were diagnosed with cervical cancer by histopathology; 3) studies that included data on first author, year of publication, ethnicity, sample size, genotype detection method, and genotype population distribution; and 4) studies published in Chinese or English.The following studies were excluded: 1) non-cervical cancer studies; 2) studies with unclear pathological diagnosis or irregular diagnostic methods; 3) studies with incomplete genotype data; 4) studies that did not report Hardy-Weinberg equilibrium (HWE) or could not calculate HWE based on reported data; and 5) reviews, animal studies, conference papers, editorials, expert opinions, letters, bioinformatic analysis articles, and case reports.The Newcastle Ottawa Scale, a tool used for assessing the risk of bias in observational studies, was used to evaluate the quality of included studies, with a total score of 9 points. Studies with a score of >5 points were considered eligible for inclusion, and those who obtained a score of >7 points were considered high quality.p value for HWE in control group.Based on a standardized strategy, two authors (Liu and Zhang) extracted the following relevant data: first author, publishing year, country, ethnicity, LncRNAs, SNP ID, sample size, genotype frequencies, genotyping methods, 2. If I2 >50% and p < 0.1, the random effects model was used to calculate the ORs and the 95% CI; by contrast, the fixed effects model was applied. Publication bias was estimated using the Egger’s test. All analyses of ORs and 95% CIs were conducted using Revman (version 5.3), and sensitivity analysis was performed using the R (version 4.0) software; a two-tailed p value of <0.05 was considered significant.The present meta-analysis was conducted based on the PRISMA checklists and followed its guideline . The oddA total of 95 literatures were obtained using our literature search strategy. After screening the titles and abstracts, 62 duplicate articles were removed, and the remaining 33 articles were used for subsequent analysis. Then, 12 articles were excluded, including 10 articles for which genotype data were not available, 1 article for which the abstract was available, and 1 article for which the full text was not available . Finallyp = 0.001; homozygote model, TT vs. CC: OR = 1.74%, 95% CI = 1.04–2.89, p = 0.03; recessive model, TT vs. TC + CC: OR = 1.84%, 95% CI = 1.10–3.07, p = 0.01) . Due to the presence of heterogeneity, a subgroup analysis was performed according to the sample size. An obvious association was detected in the allele, homozygote, dominant, and recessive genetic models. However, except for the allele model, the heterogeneity in the other four genetic models was still high. The result of sensitivity analysis showed no changes after deleting each study  . No publ = 0.01)  and was ch study .p = 0.01; dominant model, GG + GA vs. AA: OR = 1.66%, 95% CI = 1.09–2.54, p = 0.02; recessive model, GG vs. GA + AA: OR = 1.31%, 95% CI = 1.14–1.51, p = 0.001). No heterogeneity was observed in all five genetic models. As shown in Egger’s test  and funnel plots. Four eligible studies with 2,896 patients and 3,335 controls were included in the present meta-analysis to detect the association between HOTAIR rs4759314 polymorphism and cervical cancer susceptibility. p = 0.99; GA vs. AA: OR = 1.00%, 95% CI = 0.60–1.67, p = 0.98; GG vs. AA: OR = 1.00, 95% CI = 0.37–2.69, p = 0.99; GG + GA vs. AA: OR = 1.00%, 95% CI = 0.45–2.21, p = 0.98; GG vs. GA + AA: OR = 1.00%, 95% CI = 0.72–1.38, p = 0.97)  . The hetEgger’s test and funnel plot, and no publication bias was found   .Three SNPs in LncRNAs were evaluated from 11 published studies, namely HOTAIR rs920778, HOTAIR rs4759314, and H19 rs217727, in relation to cervical cancer susceptibility. These three SNPs were closely related to cervical cancer susceptibility. In the allele model, homozygote model, and recessive model, rs920778 showed a significant correlation with cervical cancer susceptibility; patients carrying T base may be more likely to develop cervical cancer. For rs4759314, in the homozygote model, dominant model, and recessive model, the G base may have a higher susceptibility to cervical cancer. H19 rs217727 showed no correlation with cervical cancer susceptibility in the five gene models. Previous meta-analyses have discussed SNPs and the risk of cervical cancer in different LncRNAs . This syPrevious studies have shown that SNPs in LncRNAs play a key role in the development of cervical cancer. LncRNA HOTAIR, encoded from the HOXC locus, is located in the region between HOXC11 and HOXC12 on chromosome 12. Previous research showed that LncRNA HOTAIR is involved in the occurrence and metastasis of cancer. The upregulation of HOTAIR expression is significantly correlated with the occurrence and metastasis of liver cancer, ovarian cancer, pancreatic cancer, neuroblastoma, and other cancer types . MultiplIn addition, the G allele of rs4759314 polymorphism can also promote the transcriptional activity of HOTAIR promoter. The G allele transfected with rs4759314 polymorphism could significantly improve the luciferase activity of cells, suggesting that the G allele of rs4759314 polymorphism could improve the RNA level of HOTAIR . Wu et aThe long non-coding RNA gene H19 (LncRNA gene H19) is located in the telomere region of chromosome 11p15.5. H19 is overexpressed in a variety of cancer types . FurtherOur meta-analysis has several advantages. First, this meta-analysis was the first to examine the association between LncRNA polymorphisms and cervical cancer susceptibility. Second, according to the methodological quality assessment, the included studies were of high quality. Furthermore, no restriction was set in the literature retrieval strategy; hence, the selection bias was well-controlled.Our meta-analysis has several limitations. First, the sample size of this study was limited, which may reduce the accuracy of the analysis. Therefore, these data need to be validated in studies with a larger sample size. Second, the search was only performed in English and Chinese databases, which may have affected the search results. If literatures in other languages are included, this study may have been more comprehensive. Finally, heterogeneity was still observed after the subgroup analysis; therefore, our conclusions should be treated with caution.In summary, rs920778 and rs4759314 polymorphisms are significantly associated with increased risk of cervical cancer, while rs217727 polymorphisms may not be associated with cervical cancer susceptibility. However, due to the small number of studies included, there is not enough data to fully confirm the association between cervical cancer and these polymorphisms; hence, the results should be interpreted with caution.


Cigarette smoking is a well-known cancer-causing behavior and a leading cause of death from cancer. However, according to previously published research and meta-analyses, cigarette smoking has a significant inverse association with prostate cancer incidence. Therefore, this study aims to investigate this association based on updated evidence by conducting a systematic review and meta-analysis.A search for relevant articles was performed in PubMed and Scopus databases to obtain the pooled relative risk (RR) and the corresponding 95% confidence intervals (CIs) for the risk of prostate cancer incidence among smokers compared to non-smokers. Our search was limited to prospective cohort studies.A total of 17 cohort studies were included in the systematic review. Fifteen studies were included in the meta-analysis and showed that cigarette smoking has an inverse association with prostate cancer incidence with a relative risk of 0.84 (95% CI: 0.78–0.91). From all cohorts included in this systematic review, five studies examined the association between current smokers and the risk of death from prostate cancer. Therefore, a meta-analysis of these cohort studies was performed and showed that current smokers had a 42% higher risk of death from prostate cancer when compared to non-smokers with a relative risk of 1.42 (95% CI: 1.20–1.68).Data from observational studies suggest that cigarette smoking has an inverse association with prostate cancer incidence. However, smokers have an increased risk of death from prostate cancer. Important to realize that this lower risk for smokers might be attributed to low prostate cancer screening uptake among smokers, misclassification bias, or selection bias from the included original studies. In summary, our results indicate that the incidence of prostate cancer is lower among smokers. Nevertheless, smokers who develop the disease have a significantly worse prognosis. The burden of PCa is predicted to grow by nearly around 2.3 million incidences and 740000 deaths worldwide by 20402. In the USA, there were an estimated 248530 new cases and 34130 deaths from PCa in the year 20213. The reported incidence of PCa in developing countries is lower than in the developed countries. It is not clear what the reason is; however, it might be due to underreporting from diagnosing centers to the national cancer registry and the geographical variation reflected by ethnic and racial dissimilarity5. Worldwide the incidence of PCa has increased remarkably, which might be attributed to the increased screening uptake among men for prostate-specific antigen (PSA) without disease symptoms6.Globally, prostate cancer (PCa) is the second leading cause of cancer and the fifth leading cause of cancer mortality among men worldwide7.Few risk factors have been identified and linked with PCa, such as a well-connected factor of racial and ethnic variation affecting the incidence of PCa. African American and Hispanic males were more likely to be diagnosed with advanced-stage PCa compared with non-Hispanic White males. Similarly, recent research demonstrated that family history and genetic factors play a role in the risk of having PCa8.Concerning age, like other types of cancer, it seems that the risk of PCa rises markedly after the age of 50 years; this might be related to the accumulated exposure to risk factors over the years. Other risk factors associated with prostate cancer but with lower evidence for a causal association, are sexually transmitted disease, unhealthy diet, inflammation, obesity, and smoking10.Regarding smoking, burning tobacco and inhaling the smoke is considered a significant risk factor or direct cause of cancer, tumor lesions, and a well-known chemical carcinogen. Yet, it is still debatable whether cigarette smoking is a causative factor of PCa12. A meta-analysis on PCa death rates among smokers showed that smoking is a modifiable environmental risk factor leading to high death rates among PCa patients. However, smoking by itself is a low-risk factor for developing prostate cancer13. Similarly, De Nunzio et al.14, reported in their systematic review that smokers have a higher mortality rate and worse outcomes after treatment.Many studies have reported that smoking and prostate cancer together can lead to a higher mortality rate, especially in heavy smokers10 conducted a meta-analysis on 24 observational studies and found that the effect of smoking by itself on PCa incidence is weak. In fact, several cohort studies showed that smoking has a significant inverse association with PCa incidence19.On the other hand, Huncharek et al.20, in their primary research published in 2014, studied 6240 men and reported that smoking was a risk factor for developing high-grade prostate cancer20. Whereas Butler et al.21 conducted a study on 27293 Chinese and Singaporean men in 2006 and found that smoking is not considered a risk factor for developing PCa. However, the same study found a slight association between those who started smoking at an early age against those who started late.In contrast, Ho et al.10, while the latest in 2014 showed an inverse association between cigarette smoking and PCa13. Therefore, this systematic review and meta-analysis aimed to assess the association between cigarette smoking and PCa risk, based on updated cohort studies.Current epidemiological evidence shows conflicting evidence about the association between cigarette smoking and risk of PCa. To the best of our knowledge, a previous systematic review and meta-analysis conducted in 2010 showed no association between smoking and PCaThe search of study of interest was done using MEDLINE and SCOPUS databases using keywords: smoking, tobacco, prostate, cancer, cigarettes, and tumor. Boolean operators (OR and AND) were used to reach published primary research that explicitly investigated this research question. The truncation * and quotation marks were used whenever appropriate to avoid missing any relevant articles when using our main keywords. In addition to keywords search, Mesh terms were exploded in PubMed to reach all articles relevant to our research question. Bibliographies were screened to conduct cross-referencing following the identification of relevant articles. Our search was limited to articles published in the English language between 2000 and October 2021. All through the search process, we followed PRISMA statement guidelines .Given that it is a systematic review and meta-analysis, our study sample was relevant published articles. Cohort study designs in the human population were included in our search. Studies that assess the risk of prostate cancer in current smokers compared to non-smokers were eligible in the meta-analysis. The minimum requirement for any study included in the meta-analysis was to that the analysis adjusted for age as a confounder. We excluded published narrative or systematic reviews, meta-analysis, case reports, letters to the editor, editorials, case-control, cross-sectional, and guidelines. The inclusion criteria of articles concerning the exposure were broad, and we accepted all types of assessments of cigarette smoking. However, the inclusion criteria for the outcome were limited to studies that used biopsy as a diagnostic tool, identification through cancer registries, and formal physician diagnosis of prostate cancer. Studies that used subjective assessment methods of occurrence of prostate cancer (e.g. by self-report directly from patients or family) were excluded.We did not attempt to contact the authors of any of the studies to request data or any additional information. The following information was extracted from each study: the first author’s surname, publication year, location (country or region) of the conducted study, follow-up time, the total number of participants and cancer cases identified among them, age of the participants, smoking categorization used, relative risk (RR) or hazard ratio (HR) estimate for each smoking category, and the corresponding 95% confidence interval, and finally the confounder variable that was adjusted for in the analysis. Data were independently extracted by two reviewers and the quality of the included studies was assessed independently as well using the New Castle Ottawa quality assessment scale (NOS) for cohort studies. A NOS score between 0 and 9 was allocated to each study. Studies with a score of ≥7 were considered high-quality studies, while studies that scored 4–6 were considered of moderate quality, and score ≤3 of poor quality. Conflicting assessments were resolved through discussion.The association between cigarette smoking and prostate cancer incidence and mortality was determined based on the relative risk (RR) and the 95% confidence interval (CIs) for each study. The hazard ratio (HR) is considered equivalent to (RR) in prospective cohort studies; therefore, all risk estimates were presented as relative risk.2 statistic, Cochran’s chi-squared (Q-test), and p<0.10 indicating significant heterogeneity. The heterogeneity points out any variability in the association between smoking and prostate cancer among diverse studies. The presence of significant heterogeneity suggests that the variation between studies is not due to chance alone. The I2 statistic ranges from 0% to 100%. The interpretation of these values falls into two categories below and above 50%, which indicate low and high heterogeneity, respectively. A sensitivity analysis was performed by excluding studies with lower quality scores and decreased follow-up time to examine the influence of such articles on the result of the crude meta-analysis. We used funnel plots for evaluating the existence of publication bias. The funnel is a scatter plot of the number of studies included in the meta-analysis; with the value of smoking effect on the horizontal axis, and the weight of the study, such as the sample size and inverse standard error, on the vertical axis. As the sample size in the studies increases, the precision in evaluating the association between smoking and prostate cancer increases. An asymmetrical pattern in the funnel plot may indicate the presence of publication bias. Statistical analyses were conducted using RevMan software (version 5.4).We calculated the pooled relative risk estimate for current smokers and the risk of prostate cancer incidence and mortality. We used a random effect model when heterogeneity was significantly high and a fixed-effect model when heterogeneity diminished. The size of heterogeneity was determined by the I23.A Preferred Reporting Item for Systematic Reviews and Meta-Analysis (PRISMA) flow diagram that summarizes the search strategy is shown in 25. Approximately the same proportion of studies was carried out in the United Kingdom27 and European countries29, while the remaining four studies were conducted in Asian countries31.A total of 2537664 participants and 47536 incident cases were included. Seven reports (42%) were from North America21. Identification of smoking categories was based on a binary measure of non-smoker versus current smokers and former smokers. However, some studies have more quantitative details for smoking, such as the number of cigarettes smoked per day, age started smoking, and pack-years smoked.All included studies in the systematic review are summarized in 30. Three studies showed a minor yet insignificant increased risk of prostate cancer in current smokers who smoke ≥20 cigarettes per day24. In contrast, the remaining two studies showed a significant inverse association between heavy smoking and the incidence of prostate cancer30.Five cohorts contained more details and statistical analysis for the number of cigarettes smoked per day in a current smoker and the risk of prostate cancer incidence compared to a non-smoker31. In contrast, the remaining twelve studies adjusted for several confounding variables such as body mass index (BMI), height, education level, family history of prostate cancer, prostate-specific antigen test, race, marital status, medication use, and other possible confounders. The last column shows the quality score for each article included in the systematic review and meta-analysis using the Newcastle Ottawa quality assessment scale for cohort studies (NOS). The vast majority of the studies scored high quality, while the rest scored moderate quality. However, one study had a possible risk of poor quality. Consequently, the study with the risk of poor quality was removed from the meta-analysis.Five studies adjusted age as a confounder in the association between cigarette smoking and prostate cancer2 statistic (68%) and (Cochran Q) p<0.0001. The pooled estimates from fifteen cohort studies that assessed the risk of prostate cancer incidence in current smokers compared to non-smokers showed an inverse association with an RR of 0.84 (95% CI: 0.78–0.91), as shown in the forest plot  and (Cochran Q) p<0.42 (Five studies examined the association between current smokers and the risk of death from prostate cancer . Therefo) p<0.42 .Funnel plots were performed to assess the potential risk of publication bias. For prostate cancer incidence meta-analysis, the shape of the funnel plots might indicate a slightly asymmetrical pattern; however, this asymmetry shows two small studies as a risk factor, and our result showed inverse association, so the potential risk of publication bias is unlikely affecting our result .Another funnel plot was performed to assess the risk of publication bias for prostate cancer mortality meta-analysis, the shape of the funnel plot showed no clear asymmetrical pattern, so the risk of publication bias is minimized .31. Similarly, another sensitivity analysis performed by excluding studies with a decreased follow-up time of 10 years or below confirmed the inverse association between smoking and the risk of prostate cancer with an RR of 0.81 (95% CI: 0.76–0.86)24. Therefore, the calculated pooled estimate in the prostate cancer incidence meta-analysis was not affected by studies with lower quality or decreased follow-up time, indicating that subgroup meta-analysis is unnecessary.A statistical analysis performed by excluding studies with (NOS) quality score equal to or below six indicating a risk of lower quality showed that smoking has inverse association with prostate cancer incidence with an RR of 0.83 (95% CI: 0.78–0.88)The pooled data in the present meta-analysis enrolling more than 2500000 participants, found that cigarette smoking has a significant inverse association with prostate cancer risk. However, a meta-analysis that included studies assessing prostate cancer mortality risk among smokers showed a 42% significantly higher risk. A sensitivity analysis was conducted due to significant heterogeny in the meta-analysis that assessed the association between cigarette smoking and prostate cancer incidence; however, neither lower study quality nor decreased follow-up period exclusion changed the crude summary effect, suggesting that the observed inverse association between smoking and prostate cancer incidence is robust and reliable.32, the association between cigarette smoking and prostate cancer incidence seems inverse. The multifactorial etiology of prostate cancer needs to be considered when interpreting these findings. Given that most casual factors in multifactorial disease have a fairly weak effect33.Although smoking is associated with various solid tumors30. The observed inverse association between smoking and prostate cancer incidence was modest. However, the association between cigarette smoking and increased mortality from prostate cancer was robust, suggesting that smoking is associated with worse prostate cancer outcomes.The result of the present meta-analysis is consistent with the previous systematic review and meta-analysis and most epidemiological observational studies that assess the risk of prostate cancer in smokers compared to non-smokers34.The search about prostate cancer mortality among smokers was not done comprehensively as this article focuses on the incidence of prostate cancer, and the data about mortality were only among the studies that measured incidence. The result was added to have a broader picture of the effect of smoking on prostate cancer. Notably, the result found in our meta-analysis complies with previous epidemiological evidence showing that smokers have higher prostate cancer mortality compared to non-smokersThis meta-analysis has several methodological strengths, including that two independent reviewers conducted the search, data extraction, and quality assessment following the PRISMA statement guidelines. Furthermore, our inclusion criteria for eligible study design were limited to cohort; therefore, the possibility of recall bias was minimized, and the clarity in temporality sequence between exposure and outcome was ascertained.In addition, the association between cigarette smoking and mortality from prostate cancer showed no heterogeneity, indicating that the risk reported in different studies was significantly consistent. We also conducted various sensitivity analyses to examine the relationship between smoking and prostate cancer in different settings.Furthermore, the funnel plots of incidence studies showed an asymmetrical pattern. Still, they were not indicative of publication bias, as small significant results showed smoking as a risk factor instead of being inversely associated with prostate cancer, as the summary estimate showed.This study has a few limitations related to the included articles that might explain the inverse association observed; most of the limitations are methodological issues related to the design or data collection tools, particularly in the assessment of smoking which might not be done comprehensively. Most studies used a smoking assessment tool at the beginning of the study with three categorizations of the exposure: none, former and current smoker. Lacking repeated exposure assessment in all of the included studies might introduce the possibility of measurement bias. Then again, this limitation will increase the risk of non-differential misclassification as the exposure might change for smokers and non-smokers equally. After all, future studies need to evaluate the exposure longitudinally and assess the intensity, duration, and intermittence of smoking to reduce the possibility of measurement bias and increase the accuracy in estimating the changing spectrum of cigarette smoking habits.31.Moreover, the results from most of the included articles were not presented comprehensively, as the data separating high-risk prostate cancer from low-risk prostate cancer were not mentioned in the retrieved studies37. This may be related to a delayed diagnosis among smokers or a stronger causal association of cigarette smoking with the prognosis of prostate cancer, which may be a plausible explanation for our findings. In either case, future studies need to consider the importance of adjusting for adherence to prostate cancer screening as a confounder.Additionally, none of the included studies mentioned any information or adjustment for prostate cancer screening. Ignorance of such an important confounder might lead to methodological bias in these studies. Important to realize that increasing evidence suggests smokers have lower compliance with prostate cancer screening programs and are more likely to have a high-grade disease at surgery, therefore a higher risk of metastasis, recurrence, and deathAnother possible explanation is that the participants who were categorized as non-smokers might be using other forms of smoking such as electronic cigarettes, hookah, or chewing tobacco, which are widely common forms of smoking. Therefore, lacking this information might explain the increased risk of prostate cancer in the non-smoker group as these details were not measured in the included articles. In order to avoid this limitation, future studies should exclude all forms of smoking in the non-smoker group when assessing the association between cigarette smoking and prostate cancer risk.1, raising global awareness about the importance of increasing smoking cessation efforts may considerably reduce prostate cancer mortality.In summary, since prostate cancer is the second leading cause of cancer and the fifth leading cause of cancer mortality among men worldwideData from observational studies suggest that cigarette smoking has an inverse association with prostate cancer incidence. However, smokers have an increased risk of death from prostate cancer. It is important to realize that this lower risk for smokers might be attributed to low prostate cancer screening uptake among smokers, misclassification bias, or selection bias from the included original studies. In summary, this result indicates that prostate cancer incidence is less among smokers, while those who smoke and develop the disease will have a significantly worse prognosis and higher mortality risk.


HF and osteoporosis shared many common etiological risk factors. However, studies exploring whether patients with HF were associated with a higher risk of osteoporotic fracture resulted in inconsistent findings. This meta-analysis aimed to summarize the association between HF and the risk of incident fracture.Following the Meta-analysis of Observational Studies in Epidemiology group recommendations, we searched multiple electronic databases  for related studies from inception to April 30, 2021. Studies evaluating the risk of incident fracture in patients with HF compared with those without HF were included for analysis. The random-effects models were used to combine the estimated hazard ratios (HRs) of incident fracture associated with HF.P < 0.001) and hip fracture . The risk of all incident fractures was increased in all subgroup analyses according to age, sample size, sex, and follow-up duration.We included 8 observational studies for meta-analysis. The sample size ranged from 5,613 to 87,748 participants, with a total of 260,410 participants included. The median follow-up duration was 5.0 years. Random-effects model analyses showed that compared with control groups, patients with HF were associated with a higher risk of all incident fractures (HR = 1.67, 95% CI = 1.30–2.16, Patients with HF were associated with a higher risk of incident fracture, as well as hip fracture. With a growing aging population world, heart failure (HF) has become a growing global public health burden. It was estimated that there were more than 37.7 million HF patients globally , 2. HF iFractures, especially osteoporotic fractures, may share some common risk factors for HF, such as aging, diabetes mellitus, cigarette smoking, and menopause , 8. ReceTo address these inconsistent data, we conducted this study, pooling data from available observational studies to summarize the association between HF and the risk of incident fracture.We performed this meta-analysis following the Meta-analysis of Observational Studies in Epidemiology (MOOSE) group recommendations  and searTwo researchers from our team independently reviewed the extracted articles and included studies for meta-analysis if they met the underlying inclusion criteria: 1. Observational studies with adult participants (aged ≥18 years); 2. Studies evaluated the risk of incident fracture in patients with HF compared with those without HF; 3. The risk of fracture associated with HF should be adjusted for confounding factors.The exclusion criteria were: 1. The association between HF and fracture was only reported in cross-sectional design; 2. Studies with a follow-up duration <1 year. If there were duplicated reports using data from the same observational study, we only included the latest published data for meta-analysis.Extraction of the data was conducted by two researchers independently. We extracted key information from the studies, including country, publication year, author, study design, study sample, mean/median age, sex proportion, HF and control cases, follow-up duration, and outcomes.We used the Newcastle–Ottawa Quality Assessment Scale (NOS) for observational cohort studies to evaluate the quality of the included studies . In brieI2 statistics were used to test the heterogeneity among the included studies, and a value of I2 > 50% was an indication of significant heterogeneity.All the statistical analyses were performed using RevMan 5.3 . The primary outcome in our study was defined as the risk of any fractures associated with HF compared with the participants. The secondary outcome was the risk of hip fracture. Multivariable adjusted hazard ratios (HRs) were extracted for pooled analysis. If the adjusted HRs were reported based on multiple different statistical models, we used the data which adjusted the most comprehensive confounders for analysis. We combined the HRs and their standard errors using the inverse variance approach. If the risk of fractures was reported as risk ratios (RRs), these data are considered approximate HRs . In caseP-values are two-tailed, and a P-value < 0.05 was considered with statistical significance.Publication bias of the primary and secondary outcomes was assessed by inspecting funnel plots. We performed the sensitivity analyses by changing the random-effects model to the fixed-effects model in the meta-analysis. Furthermore, the pooled HRs were recalculated by excluding one study at a time to determine the effects of each study on the results. We also performed subgroup analyses of the primary outcome according to sex, age, sample size, and follow-up duration. All The flow of study selection for meta-analysis was presented in I2 = 95%, P < 0.001). Random-effects model analyses showed that compared with control groups, patients with HF were associated with a higher risk of incident fractures in follow-up duration . Random-effects model analyses showed that compared with control groups, patients with HF were associated with a higher risk of hip fracture  and hip fracture  were higher in patients diagnosed with HF compared with that in controls . HoweverThere are some indications for clinical practice from our study. Both HF and osteoporotic fractures are highly prevalent in the aging society. They are common causes of disability, hospitalizations, and mortality, causing a heavy public burden to the health system and society. However, current academic guidelines for the management of treatment of HF do not propose clearly suggestions for screening, prevention and management of osteoporosis or osteoporotic fracture in patients with HF , 33. BasSeveral major strengths should be noted in our study. First, we only included studies which had adjusted for multiple confounding factors for analysis. Therefore, the association between HF and the increased risk of fractures could not be attributed to the effect of potential confounders or shared common risk factors. Second, although there was significant heterogeneity observed among the included studies for the association between baseline HF and risk of any fractures, the risk of fractures was increased in all subgroup analyses according to age, sample size, sex, and follow-up duration, which further supported the conclusion in our study. However, several limitations in the current study should be acknowledged. First, due to the observational design of the included studies, we cannot establish a causation relationship between HF and fractures. Furthermore, 7 of the included studies were from Europe or USA, and only one study was from China. Therefore, the association between HF and risk of fracture could not be generalized to other populations. Second, different medicines for the treatment of HF may have different effects on osteoporosis, and further affect the risk of fractures. However, the treatment strategies in these HF cohorts were not adjusted. Third, some important risk factors for fractures were not evaluated and adjusted, such as levels of 25 hydroxy vitamin D and bone mineral density. Only one study reported the adjustment of anti-osteoporosis medication . TherefoPatients with HF were associated with a higher risk of incident fracture, as well as hip fracture during follow-up. Better attention should be paid to ameliorating the risk of osteoporotic fracture in the management of HF.The original contributions presented in the study are included in the article/X-pL, X-yJ, J-dW, and Y-QL: research idea and study design. X-pL, X-yJ, J-dW, and D-lL: data acquisition. J-dW and J-xW: data analysis/interpretation. X-PL, X-yJ, and J-dW: statistical analysis. J-dW and Y-QL: supervision and mentorship. All authors contributed important intellectual content during manuscript drafting or revision and accept accountability for the overall work by ensuring that questions pertaining to the accuracy or integrity of any portion of the work are appropriately investigated and resolved.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.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


The result of the dose–response analysis showed that the risk of sarcopenia increased by 1.22 times for each 1-point increase in the DII score . Conclusion: The meta-analysis demonstrated that the DII is associated with sarcopenia. Considering some limitations in this study, more studies are needed to verify this relationship.Background: The dietary inflammatory index (DII) is thought to be related to many healthy events. However, the association between the DII and sarcopenia remains unclear. Methods: The meta-analysis was conducted to evaluate the effects of the DII on the risk of sarcopenia utilizing available studies. Up to September 2022, Cochrane, PubMed, Web of Science, Medline, and EMBASE databases were searched to evaluate the relationships between the DII and sarcopenia. A random‒effects model was used to calculate the effect size and 95% confidence intervals (CIs). Result: Eleven studies with 19,954 participants were included in our meta-analysis. The results indicated that a high DII increased the risk of sarcopenia (OR = 1.16, 95%CI ,  Sarcopenia is induced by a continuous loss in muscle strength, muscle function, and skeletal muscle mass . RelevanAs an important part of life, diet is closely associated with the occurrence, development, and rehabilitation of many diseases by regulating the inflammatory state . The DIIA healthy diet greatly contributes to the prevention of sarcopenia. However, in the past studies on diet and sarcopenia, people more tended to explore the intake of certain high protein foods and some trace elements. The role of the macro diet pattern on sarcopenia has been ignored. In recent years, the DII as a dietary indicator contained a variety of nutrients and reflected individual dietary patterns. Recently, more and more studies showed that the DII was related to many diseases, such as cardiovascular and cerebrovascular disease, depression, diabetes, cancer, and metabolic syndrome. Similarly, it was reported that there was a relationship between the DII and sarcopenia, and individuals with a higher DII were more likely to develop sarcopenia. However, other studies demonstrated that there was no significant relationship. The effect of the DII on sarcopenia remains unclear, and the relationship has not been demonstrated in a meta-analysis. Therefore, we conducted a meta-analysis of all related observational studies to examine the relationships between the DII and sarcopenia.The meta-analysis was carried out in strict accordance with the PRISMA guidelines, and the specific details are shown The inclusion criteria were as follows: (1) the study design was observational studies; (2) the DII was obtained based on a standard calculation method; (3) sarcopenia was defined by local criterion; (4) the literature was based on human subjects; (5) the primary outcomes of interest were presented as odds ratios (ORs) or relative risks (RRs), along with their 95% confidence intervals (CIs), or comprehensive data were provided to calculate them; and (6) duplicate articles, reviews, or conference papers were excluded. Two researchers independently evaluated all of the relevant papers, extracted the final eligible literature, discussed and resolved disagreements with experts in the field .We extracted crucial data from the final included research: first author, publication year, region, sample size, sex, mean age of participants, study design, adjusted confounding factors, performance of sarcopenia, DII in different groups, adjusted total effect estimates of sarcopenia, and their corresponding 95% CIs.The bias risk of the observational literature was assessed independently by two investigators by using the Newcastle Ottawa scale (NOS) . This scI2 statistic. A significant difference was considered to exist among the included studies if the p value was < 0.05. The degree of heterogeneity was evaluated according to the I2 value, in which 25%, 50%, and 75% indicated low, moderate, and high heterogeneity, respectively.All statistical analyses were carried out in RevMan version 5.3 and Stata version 12.0 software. The multivariate adjusted effect estimates from all of the included studies were extracted to calculate the pooled estimates and their corresponding 95% CIs via a random effects model. Potential publication bias was evaluated by funnel plot symmetry and Egger’s test. The trim-and-fill method was conducted to correct the result of bias and evaluate the effect of publication bias on the pooled estimates. A statistical heterogeneity analysis was performed by Cochran’s Q statistic and the The sources of heterogeneity were explored via a sensitive analysis and subgroup analyses. Sensitive analysis explored whether extreme results affect the pooled risk estimates by excluding each study one by one from all studies. Subgroup analyses were carried out based on sex , performance of sarcopenia  , region A total of 10,612 relevant articles were identified and initially screened from electronic databases according to the search strategy; only 11 articles  met our inclusion criteria. The specific details of included studies are summarized in I2 = 53.4%, p < 0.05, p = 0.006, p < 0.05, number of trim and fill = 4), thus indicating that the publication bias had little influence on the above results.The meta-analysis indicated that individuals with a higher DII were more likely to develop sarcopenia compared to a lower DII group  but not in the muscle strength group or physical performance group , with the risk of sarcopenia increasing by 1.22 times for each 1-point increase in the DII score  was included in the article, and the sample sizes of the contained studies varied widely. Therefore, it is necessary to conduct large-scale prospective cohort studies and basic research to obtain more conclusive evidence.In summary, the meta-analysis indicated that the DII is associated with sarcopenia. Meanwhile, the result of the dose–response analysis showed that sarcopenia increased by 1.22 times for each 1-point increase in the DII score. In light of the limitations observed in this meta-analysis, more studies are needed to verify the relationship.


The global prevalence of chronic obstructive pulmonary disease (COPD) is increasing, and the risk of lung cancer in these patients is high. The use of inhaled corticosteroids (ICSs) in COPD patients could help to decrease potential lung cancer risk. We planned to conduct this systematic review and meta-analysis to determine the role of ICS in the risk of lung cancer among COPD patients.  A comprehensive search of PubMed, Science Direct, Google Scholar, and Cochrane library and a manual search of the list of references were conducted. Studies with cohort, case-control, and randomized clinical trial designs for any ICS use reporting the incidence/hazard ratio (HR) of lung cancer were included. The random-effects model was used to pool hazard ratios. Subgroup analysis and metaregression analysis were employed. Funnel plot and Egger regression test were used to assess publication bias. p value ≤ 0.001. The use of ICS in COPD patients showed a 31% reduction in the risk of lung cancer. Subgroup meta-analysis showed a significant reduction in the risk of lung cancer as well.  Combining the results of 14 observations, the pooled HR for cancer risk reduction was 0.69 (95% CI 0.59-0.79),  The use of ICS in COPD patients reduces the risk of lung cancer. The risk reduction was independent of smoking status and latency period. Future studies should focus on the optimum dose and controlling confounders like asthma. With uneven distribution around the world, the prevalence of chronic obstructive pulmonary disease (COPD) is increasing . Over a A consistent positive association between COPD and lung cancer is common. The annual incidence of lung cancer was 2 to 4-fold higher in patients with prior COPD , 6. Fromβ (IL-1β), IL-10, IL-17A, cyclooxygenase 2 (COX 2), and T-helper 1 (Th 1) cells. The resultant effects impair cytotoxic T cell effector function, activate macrophages, and degranulate neutrophils [Mortality from lung cancer is higher in inflammatory lung conditions. Forced expiratory volume in the first second (FEV1) and C-reactive protein (CRP) significantly predict mortality from lung cancer with hazard ratio (HR) = 2.13 for FEV1 < 90% and HR = 3.38 for CRP > 2 mg/dl [trophils , 15. Thetrophils .Pharmacologic treatments for COPD include decreasing airway smooth muscle contraction, airway inflammation, mucus production, respiratory infection, and replacement of alpha-1 antitrypsin deficiency. Inflammatory conditions in COPD can be managed with corticosteroids, phosphodiesterase inhibitors, and long-term macrolides . In the ICS use for COPD has mixed outcomes; deleterious effects like pneumonia  and tubeAs the global prevalence of COPD is increasing and the risk of lung cancer in these patients is high; the use of ICS in COPD patients could help to decrease potential lung cancer risk. A systematic review by Raymakers et al.  in 2016 Our inclusion criteria were as follows: (1) studies with a cohort, case-control, and randomized clinical trial designs for any inhaled corticosteroids including beclomethasone, budesonide, triamcinolone, fluticasone, ciclesonide, mometasone, flunisolide; (2) the follow-up period should be at least 12 months; (3) study participants with COPD; (4) the intervention should be ICS, the control group with placebo or other drugs with a specified name; and (5) data on the incidence of lung cancer including zero events in the course of follow-up.In vitro and animal studies, review articles, incomplete articles, conference proceedings, and duplicates were excluded. Due to feasibility, articles published in the English language/have English version were included. All the abovementioned designed research conducted on adults up to November 2021 were included.PubMed, Science Direct, Google Scholar, and Cochrane library searched for the following terms without publication year restrictions: COPD, “obstructive pulmonary disease,” “Pulmonary Disease, Chronic Obstructive,” “inhaled corticosteroids,” Beclomethasone, budesonide, triamcinolone, fluticasone, mometasone, ciclesonide, flunisolide, lung neoplasms, lung cancer, non-small lung carcinoma, small lung carcinoma, adenocarcinoma, “carcinoma, squamous cell” search words used alone or in combination using Boolean operators. Unpublished articles were searched from clinical trial registration platforms. Preprint articles were also retrieved from websites. A manual search was conducted by screening the reference lists of included studies. Reports were compiled according to the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) guideline .Two reviewers independently scanned all titles and abstracts that indicated the use of corticosteroids by a patient with COPD and the risk of lung cancer. After obtaining full reports of potentially relevant studies, the same reviewers independently assessed the full-text articles. Disagreements regarding eligibility were solved by the third reviewer through consensus.General study details including author, publication year, study design, country, follow-up period and duration, age of participants, the sample size in both groups, types of treatment, latency period, adjusted variables, and hazard ratio were extracted from each study with predetermined data abstraction format. The outcome of interest was the risk of lung cancer, incidence, or HR. In studies that had more than two groups and reported more than one hazard ratio, the HRs will be taken accordingly. In this case, the sample size of the control group was divided by the number of groups/experimental arms to avoid double counting.The methodological quality of observational studies was assessed using the Newcastle Ottawa Scale (NOS) . The risI2 static, and we consider percentages of around I2 = 25%, I2 = 50%, and I2 = 75% as low, medium, and high heterogeneity, respectively [We performed a meta-analysis to pool HR with a 95% confidence interval (CI) from dichotomous data using random effects, inverse variance method. Statistical heterogeneity was measured by ectively .p value ≤ 0.05 cut-point was used to declare statistical significance. To detect the robustness of the results, a sensitivity analysis was conducted by sequential elimination of each study from the pool. All analyses were performed using the STATA software .Subgroup meta-analysis was performed for potential sources of clinical and methodological heterogeneity. Studies subgrouped based on the status of adjustment for smoking, duration of follow-up period, consideration of time bias, and so on. Meta-regression analysis was performed for potential covariates. Potential publication bias was assessed using a funnel plot and Egger's regression test. A total of 971 potentially eligible studies identified from the four databases were searched, and three additional articles were retrieved from other sources. Two hundred one duplicates were removed with 773 studies left to deal. After screening the titles and abstracts of these studies, 742 papers discarded as irrelevant. Thirty-one studies selected for full content screening, and 13 articles met the inclusion criteria. The detail of the literature selection is given in μg/day without shared HR. We took these two HRs as two observations, and the total observations are 14 with 13 unique studies. Nine cohort studies, three case-control, and one randomized controlled trial, a total of 13 studies, (with 14 observations), were included. The mean follow-up time ranges from 3 to 6.8 years, and publication year ranges from 1999 to 2019. Parimon et al.  reportedp value ≤ 0.001. The use of ICS in COPD patients showed a 31% reduction in the risk of lung cancer with a mean follow-up time of 3-6.8 years. However, there is borderline moderate heterogeneity I2 = 61.2%. Subgroup meta-analysis was conducted on relevant factors including study design, continent, smoking status, and latency period. All factors fail to change the significant reduction in the risk of lung cancer among COPD patients taking ICS  and not adjusted . In both cases, there is a significant risk reduction, and adjustment for smoking resulted in a greater reduction of 36% (vs. 28%). Subgroup analysis based on adjustment for smoking is given in Cohort studies reported greater significant reduction in risk, than case control studies, 32% and 26%, respectively, with moderate heterogeneity in cohort studies. A single RCT included in this analysis insignificant result. Continent wise, researches conducted in Europe had pronounced risk reduction than elsewhere. There was also significant risk reduction in Asia and America (USA and Canada) with 28% and 23%, respectively, although lower than the European studies.p value ≤ 0.001. Without latency period, risk reduction was higher (HR = 0.61 (95% CI 0.47-0.76), p value ≤ 0.001). Forest plots for subgroup meta-analysis based on study deign, continent of the study, and latency period are shown in the supplementary material, supplementary figures The latency period is the period between the entrance to follow-up and the development of lung cancer. A one-year latency period was allowed in some studies, and we performed a subgroup analysis to test if the latency period had affected the outcome. As we can see in We also conducted a metaregression analysis to identify covariates affecting the outcome. Metaregression was undertaken on study design, continent, smoking adjustment, and latency period; the result showed none of them were significant covariates. Metaregression analysis results are given in p value = 0.476). Sensitivity analysis was performed by sequential elimination of each study from the pool, and there was single study affecting the finding, indicated in the supplementary material  significant risk reduction. The evidence was synthesized from observational and interventional studies . Although it is well established that COPD is an independent risk factor for lung cancer, the mechanism by which ICS use could reduce the risk is not well understood and more studies are being conducted.Chronic inflammation is in the first line of proposed mechanisms. Thomsen et al.  measuredThe capability of inflammatory cells to produce arachidonic acid metabolites was decreased after fluticasone inhalational treatment ; inhalatThis meta-analysis showed ICS treatment has reduced the risk of lung cancer. The risk reduction was independent of study design , smoking adjustment, or latency period. In observational studies, both cohort and case-control, a significant risk reduction was observed. However, the results of RCTs were not pooled because of insufficient eligible studies. Adjustment for smoking still provided significant risk reduction, but in comparison with unadjusted risk, the risk reduction was greater.Incorporating a latency period in the analysis of primary studies was an important component of reducing time-related bias. An incident case that occurred within days of the entrance to follow-up will not indicate the effect of treatment. Since lung cancer is diagnosed at the advanced stage, it is likely to have been present prior to the clinical diagnosis. This has been the source of variation in different reports. Given the biology of tumor growth in lung cancer, a year latency period was applied in some studies . SubgrouAlthough this study provided comprehensive evidence from good quality literature (quality of studies indicated tables The use of ICS in COPD patients reduced the risk of lung cancer. The risk reduction was independent of smoking status and latency period. Future studies should focus on the optimum dose and controlling confounders like asthma.