MUC5B promoter rs35705950 T/G gene polymorphism has been associated with the risk of IPF, but the influence of this relationship varies among different populations. In the past 2 years, there were new clinical studies with different results, but none of them reached unified conclusions. Therefore, this study further included the latest case–control studies, integrated their results and carried out meta-analysis on them to draw reliable conclusions. PubMed, EMBASE, CNKI, Wanfang database and VIP Chinese science were searched by a computer to collect the related literatures of MUC5B gene polymorphism and IPF susceptibility published before June 15, 2021. The first author, year of publication, diagnostic criteria and gene frequency were extracted after screened them. Forest plot was drawn and the trial sequential analysis (TSA) was carried out to confirm the stability of the meta-analysis results. Registration number: CRD42021272940. A total of 24 case–control studies , and a total of 6749 IPF patients and 13,898 healthy controls were included in this study. The T vs.G, TT vs. GG, GT vs. GG, GT + TT vs. GG and TT vs. GG + GT genetic models of MUC5B promoter rs35705950 T/G polymorphism were associated with IPF risk in all populations, and the effect values were , , ,  and , respectively. The results of TSA confirmed the stability of the results. Subgroup analysis showed that T vs.G, TT vs. GG, GT vs. GG, GT + TT vs. GG and TT vs. GG + GT genetic models of MUC5B polymorphism were associated with IPF risk in Caucasian population. The effect values were , , ,  and , respectively. Similar results were also found in Asian and mixed populations. The association strength of the minor T allele in the Caucasian was more significant than that of the Asian population ([OR] 4.50 vs. [OR] 2.39), and the association strength of all genetic models carrying "T" was more significant than that of the Asian population ([OR] 10.98 vs. [OR] 4.29). In Caucasian, Asian and mixed populations, T minor allele carriers were more likely to be susceptible to pulmonary fibrosis, and TT genotype carriers were more likely to be susceptible to IPF than GT genotype carriers. The association between IPF and Caucasian population with minor T allele and all "T" genetic model was more significant than that of Asian population. In recent years, the incidence of IPF has been increasing, and the prevalence increased from 276 cases per 100,000 in 2010 to 725 cases per 100,000 in 20193. At present, the cause of IPF is not yet clear, but the role of genetic factors has received more and more attention in the studies of its pathogenesis. A study had shown that as many as one fifth of patients affected by IPF reported that their family members had pulmonary fibrosis4. Mucin 5B (MUC5B) played an important role in immune regulation in maintaining bronchoalveolar epithelial function, and its genetic variation had been identified as a risk factor for IPF6.So far, Genome-wide Association Study (GWAS) has found a SNP in the promoter region of MUC5B gene (rs35705950), in which the T allele frequency was 30–35% in IPF cases14.Variant rs35705950 alone explains 5.9–9.4% of disease liability in the general population and 13.5% in people > 65 years of age15. MUC5B promoter SNP rs35705950 and its enhanced expression in distal airway epithelial cells are considered to be related to the pathogenesis of IPF19. Due to excessive lung injury and abnormal repair, excessive release of MUC5B can lead to the exacerbation of IPF22.SNP rs35705950 is a G to T trans version in the 5 'flanking region of MUC5B 3 kb upstream of the transcription initiation site, it shows the strongest association with IPF14.In the Caucasian population, 41.9% of IPF patients and 10.8% of the control group had the T-risk allele11, while in the Chinese population, the frequency of T allele was about 3.33% in IPF patients and 0.66% in the control group23. In recent years, meta-analysis has showed that comparing with G allele, minor T allele was associated with increased risk of IPF, TT genotype carriers were more prone to IPF than GT genotype carriers, and the association strength of Caucasian population was more significant than that of Asian population26.However, in the past two years, there were new clinical studies reporting the correlation between rs35705950 and IPF, and their research results were different, so there was no unified statement.Idiopathic pulmonary fibrosis (IPF) was a disease with insidious onset, short course and high mortalityTherefore, based on the results of previous clinical studies, this study further included the latest case–control studies, integrated their results and carried out meta-analysis on them to draw reliable conclusions.https://www.crd.york.ac.uk/prospero/). The procedure of this protocol is based on PRISMA-P guidance27.This study has been registered in PROSPERO, registration number: CRD 42021272940  The gene frequency data is complete and can be used to calculate the OR and 95% CI; (4) The distribution of genotype frequency of all subjects conformed to Hardy–Weinberg equilibrium28; (5) The score of Newcastle Ottawa scale (NOS)29 was no less than 7(≧7).(1) The case–control studies are all based on the susceptibility of MUC5B rs35705950 T/G gene polymorphism and IPF; the language is either Chinese or English; the description of detection methods and means is accurate; (2) They conform to the authoritative standards established by the Chinese Society of Respiratory MedicineConference reports, reviews, case reports, the study failing to obtain allele frequency and research based on pedigree were not included. The same study published many times, only the one with the largest sample size and the most complete information was reserved.The pre-specified primary outcomes were to investigate whether MUC5B T/G increased the risk of IPF in the entire population. The secondary outcomes were to determine whether there was a difference in the strength of the association between MUC5B T/G and IPF among different ethnic groups.PubMed, EMbase, CNKI, WanFang Database, VIP were searched through the Internet to collect domestic and foreign related literature data published before June 15, 2021on MUC5B gene polymorphism and IPF susceptibility. Theme words and keywords were retrieved combining with literature retrospective and manual retrieval methods. Search terms: "rs35705950" or"MUC5B"and"polymorphism"and"Idiopathic pulmonary fibrosis" or "IPF". The languages were limited to Chinese and English. Table Two relatively independent researchers (X–Z W and W L) conducted literature screening and data extraction. After excluding the studies that obviously did not meet the inclusion criteria, they further read the full text of the studies that might meet the inclusion criteria to determine whether they could really be included, and then they cross checked them. They discussed to resolve or submitted to the third party (Y–Z C) for ruling when there were different opinions. If the report was not clear or lack of information, they tried to contact the author of the original text by e-mail to further obtain relevant data. The extracted data include: general clinical data, research subjects, year of publication, country of research subjects, race of research subjects, diagnostic criteria of IPF, number of cases in case group and healthy control group, and frequency of each genotype.29. Quality score ranged from 0 to 9, and the score of more than 7 were considered as high-quality ones.Two independent researchers (X–Z W and W L) evaluated the selected literatures according to NOShttps://training.cochrane.org/ and stata12.0 (Stata/SE 12.0) for Windows (32-bit). Revision 22Apr 2015.Copyright 1985–2015 StataCorp LP. URL link: http://www.stata.com software were used to process data; Hardy–Weinberg genetic balance of subjects was analyzed by Pearson test; the heterogeneity between studies and subgroups was evaluated by Q-test and I2. If P < 0.1 or I2 > 50%, it was considered that there was heterogeneity between studies. The data were combined by random effect model, and OR and 95% CI were calculated; If there was no heterogeneity between studies and subgroups, the fixed effect model would be used for data consolidation; The OR value was calculated according to the data of allele genetic model (T vs. G), dominant genetic model (GT + TT vs. GG), recessive genetic model (TT vs. GG + GT), additive genetic model (TT vs. GG) and heterozygous genetic model (GT vs. GG);The forest plot was drawn to show the research results and their characteristics; Publication bias was judged by funnel plot, and evaluated by Begg's test and Egger's test. Sensitivity analysis was performed for the results with high heterogeneity. With reference to previous studies31, we used TSA software . Version 0.9.5.10 Beta. Copenhagen: Copenhagen Trial Unit, 2016. URL link: https://ctu.dk/tsa/ to perform TSA tests to evaluate the stability of the conclusion (Type I error) probability = 5%, statistical test power = 80%, relative risk reduction = 20%).Revman5.3 (Review Manager (RevMan)Version 5.3. Copenhagen: The Nordic Cochrane Centre, The Cochrane Collaboration, 2014. URL link: This review does not require ethical approval because the included studies are published data and do not involve the patients’ privacy. The results of this review will be reported in accordance with the PRISMA extension statement and disseminated to a peer-reviewed journal.10 was not included as a broad case definition was used (fibrotic IIP) and IPF specific results were not reported, meaning this paper did not meet our inclusion criteria. Figure 27, and Table The five databases were first detected with 198 related literatures, and after a layer by layer screening, 24 studies were finally included, including 13 studies in Caucasian, 7 in Asian and 4 in mixed population. A total of 6749 IPF patients and 13,898 healthy controls were included, including 5100 Caucasian, 1090 Asian and 559 mixed population patients. Fingerlin et al.24 trials in this study were evaluated by NOS scale. Except for the 2 trials that were moderate (scores = 7), all other trials had high results (scores = 8), indicating that the risk of bias was relatively low. The results are shown in Table 2 = 60%, and the T minor allele was associated with the risk of IPF compared with the G allele  , Asian , mixed , each subgroup   , Asian , each subgroup , there was no heterogeneity test result  , Asian , each subgroup , there was no heterogeneity test result  , Asian , each subgroup , there was no heterogeneity test result  , Asian , each subgroup , there was no heterogeneity test result  was also one of the most important genetic associations with IPF.GWAS found that the minor T allele of single nucleotide polymorphism (SNP) rs35705950 was present at a frequency of 38% in European subjects with IPF37. It has been found that the expression of MUC5B in subjects with IPF was 14.1 times higher than that in subjects without IPF14. Excessive production of MUC5B led to IPF due to excessive lung injury and abnormal repair54. Therefore, in recent 10 years, people have been studying the relationship between this gene polymorphism and IPF susceptibility. At present, some related studies have been published, but no consistent conclusion has been reached. Based on the meta-analysis, the data with the same research purpose can be summarized and analyzed, which increases the effectiveness of the test so as to draw more objective and reliable conclusions. Therefore, this study conducted a meta-analysis on the relationship between the polymorphism of MUC5B rs35705950 gene and the susceptibility to IPF.The incidence of Idiopathic pulmonary fibrosis (IPF) has been increasing year by yearA total of 24 case–control studies with 6749 IPF patients and 13,898 healthy controls were included. The genotype and allele distribution frequencies of all subjects were in accordance with Hardy Weinberg equilibrium test, which showed that the selection of subjects was representative of the population, and the samples were in a random distribution equilibrium and a wide range of population. The results showed that rs35705950 T/G polymorphism of MUC5B promoter was associated with IPF risk in T vs.G, TT vs. GG, GT vs. GG, GT + TT vs. GG and TT vs. GG + GT genetic models, meanwhile, sensitivity analysis and publication bias analysis showed that the results were stable and reliable. The results of test sequential analysis (TSA) also confirmed the stability of the results. These showed that people carrying the T minor allele were more likely to be susceptible to pulmonary fibrosis, and the T minor allele was a risk factor for the onset of pulmonary fibrosis. It was found that carriers of TT genotype ([OR] 10.12) were more likely to develop IPF than carriers of GT genotype ([OR] 4.84) when comparing the OR values of different genetic models.Considering ethnic factors may have an impact on the results, we conducted a subgroup analysis: In the Caucasian population, the MUC5B gene polymorphism was in T vs. G, TT vs. GG, GT vs. GG, GT + TT vs. GG and TT vs. GG + GT genetic models were related to the risk of IPF, and the results of TSA and publication bias analysis confirmed the stability of the results. These showed that Caucasian populations carrying T minor alleles were more susceptible to pulmonary fibrosis, and T minor alleles were a risk factor for the onset of pulmonary fibrosis in the Caucasian population. TT genotype carriers ([OR] 10.98) were more likely to be susceptible to pulmonary fibrosis than GT genotype carriers ([OR] 6.27). Similar results were also found in Asian and mixed populations. It showed that Asian populations and mixed populations carrying T minor alleles were also susceptible to pulmonary fibrosis. T minor alleles were also the risk factors for the onset of pulmonary fibrosis in Asian populations and mixed populations. TT genotype carriers were more susceptible to IPF than GT genotype carriers . For the Caucasian population and the Asian population, the association strength of the minor T allele in the Caucasian was more significant than that of the Asian population ([OR] 4.50 vs. [OR] 2.39), and the association strength of all genetic models carrying "T" was more significant than that of the Asian population ([OR] 10.98 vs. [OR] 4.29).33: A recent retrospective study reported the susceptibility of 62 Caucasian IPF patients who were followed up from 2012 to 2019. The results showed that the MUC5B rs35705950 minor T allele was more common in IPF subjects than in healthy subjects . In addition, Stock et al.43 observed increased expression of MUC5B in T allele carriers of Caucasian IPF patients (n = 23). A recent case–control study42 in Chinese Han population showed that the frequency of GT + TT genotype and T allele in patients was significantly higher than that in controls. The OR of IPF in T allele carriers was 2.603, 95% CI was (1.268–5.343). The average overall survival (OS) of patients with GT and TT genotypes of MUC5B rs35705950 was significantly lower than that of patients with GG. It is speculated that the polymorphism of MUC5B rs35705950 gene may be a risk factor for IPF in Chinese Han population, and the polymorphism of MUC5B rs35705950 gene was related to the decreased mortality of IPF patients.The above conclusions were very similar to the results of clinical studies in the last yearHeterogeneity was generally considered to be the main factor affecting the reliability of meta-analysis results. In this study, it’s confirmed that there was heterogeneity except TT vs. GG + GT genetic model, and there was also some heterogeneity between subgroups. The heterogeneity of Caucasian population was obvious, but after sensitivity analysis, it was found that the results of each genetic model were relatively stable. Further publication bias analysis did not find any bias, indicating that the results of this study were generally stable and reliable. In terms of sample size, although the sample size of each genetic model in all populations basically met or approached the sample requirements of TSA. However, after subgroup analysis, the sample size of each population still did not meet the requirements of TSA, especially in Asian and mixed groups, which will affect the credibility of the results of Asian population and mixed population to a certain extent. Although the TSA results confirmed the stability of the T vs. G results of Asian and mixed groups, there was a certain publication bias in these results. The reason for this publication bias may be the relatively small sample size. In addition, for other genotypes in Asian populations and mixed populations, the relatively small sample size made it impossible to judge the stability and publication bias of these results. Therefore, for the results of Asian populations and mixed populations, more samples will be needed to confirm it. In addition, because the original studies included did not explore the content of gene–gene and gene–environment, and couldn’t collect relevant data, the interaction between environment and gene could not be further analyzed, which all caused the limitations of this meta-analysis.The meta-analysis overcame the difficulties mentioned above and reflected the relationship between IPF and different ethnic groups as much as possible. The meta-analysis results showed that people with T minor allele in Caucasian, Asian and mixed populations were more likely to be susceptible to pulmonary fibrosis, and those with TT genotype were more likely to be susceptible to IPF than those with GT genotype, moreover, the association strength of minor T allele in Caucasian population was more significant than that in Asian population, and the association strength of all genetic models carrying "T" was more significant than that in Asian population. But of course, more and higher quality large sample case–control studies will be needed to verify the results in Asian population to provide more effective basis for the prevention and treatment of IPF.Supplementary Information.


In ethnicity subgroup analysis, we found that the TNF-α -308G/A polymorphism was associated to COPD among Asians  when smoking status was not adjusted. However, no significant association was found in Asian smokers or Caucasian smokers. In conclusion, our study suggest that TNF-α-308 GA genotype is related to COPD in the Asian population. In addition, the TNF-α+489G/A, - 238G/A variants do not increase the risk of COPD.The tumor necrosis factor alpha (TNF-α) polymorphism may play an important role in chronic obstructive pulmonary disease (COPD) susceptibility. However, the results are still inconclusive. Eligible studies were searched in Cochrane Library database, EMBASE, Pudmed, Web of science, China National Knowledge Infrastructure, and Wanfang database. Finally, a total of 27 case-control studies with 3473 COPD cases and 4935 controls were included in the present analysis. We also performed trial sequential analysis (TSA) to confirm our results. Overall, association between TNF-α-308G/A polymorphism and COPD susceptibility was identified in allelic model (A vs. G, OR = 1.21, 95%CI: 1.01–1.45, Systematic Review Registration: https://www.crd.york.ac.uk/PROSPERO/, identifier CRD42021273980. Chronic obstructive pulmonary disease (COPD) is characterized by progressive airflow obstruction and bronchial hyperresponsiveness, with increasing morbidity, mortality, and resource utilization worldwide . AccordiTumor necrosis factor alpha (TNF-α) gene is found on chromosome 6p21.33 . The TNFWe followed the PRISMA guidelines and registered the review protocol on PROSPERO .A comprehensive search of PubMed, EMBASE, Web of science, Cochrane Library database, Wanfang databases, and China National Knowledge Infrastructure were performed to find eligible studies. We retrieved articles using search strategies: “TNF-α or Tumor Necrosis Factor-α” and “COPD or COAD or chronic obstructive lung disease or chronic obstructive pulmonary disease or chronic obstructive airway disease” and “polymorphism or SNP or genotype or variant” published before July 25, 2021. In addition, we reviewed all references cited in the identified articles.The included studies had to meet all four criteria: : Case-coTwo investigators retrieved data from the included studies separately, and all discordances were discussed to reach an agreement. The extraction information mainly includes the first author’s name, year of publication, country of origin, ethnicity, source of control, genotype method, the overall number of cases and controls, and genotype distribution of three TNF-α gene variants. The Newcastle-Ottawa Scale (NOS) was applied to estimate the quality of included articles, and the studies were classified as high quality (scores ≥7 stars) or low quality (scores<7 stars).p > 0.1 and I2<50%) indicated no heterogeneity among studies, and fixed-effects model (the Mantel-Haenszel method) was applied. Otherwise, when I2 ≥ 50% or p ≤ 0.10, we performed random-effects model (the DerSimonian and Laird method). Publication bias of the literature was evaluated by funnel plots and Egger’s test, and one-way sensitivity analyses were performed to assess the stability of the results.The odds ratio (OR) and its 95% confidence interval (95%CI) were used as an effect size to assess the risk of COPD caused by TNF-α polymorphism, as well as the strength of association between them. Z-test was used to estimate the statistical significance of pooled ORs. There were five genetic comparison models: allelic model (A vs. G), heterozygote model (GA vs. GG), homozygote model (AA vs. GG), dominant model (GA + AA vs. GG), and recessive model (AA vs. GG + GA). We checked the heterogeneity assumption by the Q-test. The outcome  and the trial sequential monitoring boundary (TSMB). The results are considered reliable and stable when the Z-value crosses TSMB. At the same time, the sample size can be deemed adequate. However, if the cumulative Z value does not cross the TSMB or RIS threshold, it means the sample size is not sufficient. And it still needs more studies to confirm the result.We used the TSA v0.9.5.10 Beta software to perform the trial sequential analysis. Our study set the relative risk reduction to 20%, the first type of error The search strategy yielded 261 potentially relevant articles, and 27 publications with 3473 COPD cases and 4935 controls were ultimately included in the present analysis . All thep = 0.04), using an allelic model. According to the comparison between heterozygote model  and dominant model , GA genotype carriers of TNF-α-308 have a higher risk of developing COPD compared to GG carriers.The effect of the TNF-α-308G/A polymorphism on COPD was first investigated by pooling 25 studies where smoking status was not adjusted, comprising 3283 COPD cases and 4539 non-COPD controls. It indicated that the A allele was associated with an increased COPD risk in the overall population . The pooled OR of heterozygote model was 1.35 , indicating that the risk of COPD with TNF-α-308 GA genotype was 1.35 times higher than that with the GG genotype in an Asian population. A similar result was also observed in the dominant model , which further supported our finding. In contrast, no significant association between the TNF-α-308 G/A SNP and COPD was found in Caucasians . Subgroups analysis by ethnicity also showed no association between TNF-α+489 and COPD among Asians (p > 0.05) and Caucasians (p > 0.05)  .p > 0.05). . .Sensitivity analysis was performed by sequentially excluding each study to assess the stability of the results in this meta-analysis. As shown in p > 0.05), and the results suggested that there was no publication bias for the association between TNF-α-308 polymorphism and COPD in included studies . The results on TNF-α-308 showed that the Z-curve crossed the TSMB . TherefoFor the TNF-α-308 G/A, the GA genotype was significantly associated with increased COPD risk, especially among Asians. Presumably, the GA variant was a stronger activator of TNF-α transcription than the GG genotype . Some stAs for the TNF-α+489G/A, no significant association was found between TNF-α+489G/A polymorphism and COPD susceptibility. Due to the new case-control study  includedTo the best of our knowledge, this is the first meta-analysis to analyze TNF-α -238G/A variant and its contribution to risk of COPD. Similarly, the pooled results suggested that TNF-α -238G/A polymorphism had no significant association with COPD risk. All the case-control studies about TNF-α-238G/A we retrieved were restricted to Caucasians. Hence, we are looking forward to more original studies on other ethnicities to confirm our conclusions.Our study features some advantages. Firstly, we performed TSA to reduce the risk of type I error and evaluate required information size. The TSA result showed that the conclusion about TNF-α-308 G/A was based on a sufficient number of cases and controls. Secondly, we carried out a second meta-analysis restricted to smokers. As smoking is the major risk factor for the development of COPD, this sub-group analysis illustrates the additional information genetic polymorphism provides when studying risk factors of COPD. Thirdly, NOS quality test was performed to estimate the quality of the case-control studies included, making the results more reliable. In addition, sensitivity analysis and Egger’s test were performed to assess the publication bias. And the test showed our results were stable in the Asian subgroup.Nevertheless, there are still some limitations that cannot be avoided in our meta-analysis. First, due to the lack of case-control studies related to nonsmokers, we failed to ascertain the association between the TNF-α-308 G/A variant and COPD risk in nonsmoking populations. Second, some unpublished studies might not be retrieved. Third, there were limited data to conduct subgroup analyses in Asians and Caucasians. The conclusions for TNF-α+489G/A and TNF-α -238G/A required more original studies to make them convincing.Together, the present meta-analysis indicated that TNF-α-308G/A polymorphism was associated with an increased risk of COPD among Asians but not in Caucasians, and GA genotype carriers of TNF-α-308 had a higher risk of developing COPD compared to GG carriers when smoking status was not adjusted. However, in the meta-analysis with restrictions to smokers, no association was found between TNF-α-308 polymorphism and COPD susceptibility either in Asian smokers or Caucasian smokers. Moreover, the TNF-α+489G/A, -238G/A variants did not have an increased risk of COPD susceptibility.


The normal reference value of the global longitudinal left atrial strain during the reservoir phase (LASr) by 3D speckle-tracking echocardiography (3DSTE) is needed to define the abnormal and normal spectra and to compare and interpret the obtained values. The present study is a meta-analysis of 3DSTE-derived normal reference value of the longitudinal LASr and an attempt to determine probable contributing factors in the variations of reported ranges. The databases of PubMed, Scopus, and Embase were searched for the following keywordS: “Left atrial/left atrium” and “strain/speckle/deformation” and "three-dimensional/3-dimensional/three dimensional/3 dimensional/three dimension/3 dimension/three-dimension/3-dimension/3D/3-D". The studies selected included those on adult healthy subjects without cardiovascular risk factors. A random-effect model was used to calculate the global 3DSTE-derived longitudinal LASr, and meta-regression was applied to determine inter-study heterogeneity. Our search yielded 316 adult subjects from 5 studies. The mean value of the global 3DSTE-derived longitudinal LASr was 27.5% . There was significant heterogeneity between the studies. The meta-regression analysis revealed the publication year, the heart rate, and systolic and diastolic blood pressure as the sources of heterogeneity. The current meta-analysis determined a normal reference value of the global 3DSTE-derived longitudinal LASr of 27.5% . The heterogeneity between studies may be explained by the publication year, the heart rate, and systolic and diastolic blood pressure. The LA wall is thin by comparison with the LV wall. The chamber regulates LV filling and works in interaction with the LV. The LA has 3 phasic functions during the cardiac cycle: reservoir, whereby blood is stored during the LV systole; conduit, whereby blood is transferred to the LV in early diastole; and contraction, whereby blood is pushed into the LV in late diastole2. The phasic functions of the LA can also be evaluated by several echocardiographic methods such as 2D or 3D volumetric parameters, pulsed-wave Doppler, tissue Doppler imaging, and myocardial deformation imaging3. Not only are the deformation markers less load-dependent2 and less affected by tethering motion, but also they are angle independent4. The use of 2D speckle-tracking echocardiography (2DSTE) has been widespread for the evaluation of the LA phasic functions, and the modality has received attention from echocardiography societies, which have presented standardization recommendations5. However, 3D deformation markers have been validated in experimental and human studies6, and deformational parameters obtained by 3D speckle-tracking echocardiography (3DSTE) have more agreement with cardiac magnetic resonance tagging than those obtained by 2DSTE7. By comparison with 2DSTE, 3DSTE is not affected by out-of-plane motion or twisting motion produced by motion in the third direction8. Moreover, 3DSTE is a feasible and reproducible method8 that has been applied to evaluate the LA phasic functions in various clinical conditions such as atrial fibrillation, hypertrophic cardiomyopathy, hypertension, and healthy subjects13.The LA phasic functions can be evaluated by several modalities such as echocardiography, computed tomography, and cardiac magnetic resonance imaging14, 3DSTE-derived normal reference values for the LA deformation indices have yet to be determined. Normal reference values are necessary for the recognition of abnormal values, and they can lead to more clinical application of the LA assessment by 3DSTE.Whereas 2DSTE-derived normal values for the LA deformation indices have already been presented15. Accordingly, in the present study, we sought to determine not only normal reference values for the longitudinal 3DLASr in adults through a systematic review and a meta-analysis but also the cause of variations in values.The measurement of the longitudinal LA strain during the reservoir phase by 3DSTE (3DLASr) or 2DSTE (2DLASr) has been done in most published articles in this field rather than strain values in other directions or phases16. The databases of PubMed, Scopus, and Embase were searched for the following keywords: “Left atrial/left atrium" and “strain/speckle/deformation” and "three-dimensional/3-dimensional/three dimensional/3 dimensional/three dimension/3 dimension/three-dimension/3-dimension/3D/3-D"  guidelinesStudies were included if they reported longitudinal LASr as numbers (the mean and the standard deviation vs only the mean in figures) in healthy subjects. Exclusion criteria consisted of animal studies, review articles, case reports, letters to editors, editorials, and conference abstracts. Also excluded were studies without control groups, studies with control groups containing healthy subjects aged below 18 years, studies featuring control group patients with cardiovascular risk factors, studies with inadequate descriptions of control groups regarding the absence of cardiovascular risk factors, and studies reporting the averaged values of the LA segments rather than the global value. This stage was done independently by 3 investigators . Disagreements were resolved according to consensus between A.H. and T.D.The eligible articles were reviewed by the three investigators  independently. Study characteristics, clinical data, echocardiographic method characteristics, echocardiography data, and 3DLASRr were extracted. The disagreements were resolved according to consensus between A.H. and T.D. Among the studies that used the same data set, the studies selected were those that (a) included control subjects not matched with a case group, (b) were not confined to a special subgroup, and (c) had more control subjects.P < 0.1) and the I2 statistic were used for the evaluation of heterogeneity between the studies. The results were illustrated as a forest plot. Clinical and demographic data, if reported by more than 2 studies, were considered a possible source of 3DLASr variations and evaluated by meta-regression to estimate their effects on the variation of the normal reference value of 3DLASr. The results were presented as β and 95% CI. The stability of the estimated normal reference value of 3DLASr was presented through a comparison between the random-effects model and a fixed-effects model by excluding a study conducted with General Electric software and by excluding the studies that had less CI overlap with other studies. Publication bias was assessed by the Egger test (P < 0.1).The statistical analysis was done by Stata software, version 16 . The mean and the 95% confidence interval (CI) of 3DLASr were computed by the random-effects model. The Cochrane’s Q test 13.Figure P < 0.001, I2 = 85.7%) . The range of the global longitudinal 3DLASr was 23.7–31.0%. The heterogeneity between the studies was significant . Following the exclusion of the General Electric software-based study, the mean value of the global longitudinal 3DLASr was 28.4%  according to the random-effects model. The Toshiba-based study forest plot is presented in Fig. P = 0.026), the heart rate , and systolic  and diastolic  blood pressure were recognized as the sources of inter-study heterogeneity .The assessment of publication bias revealed no significant publication bias for the global longitudinal 3DLASr , which is more than the value we obtained for 3DLASr  . The aforementioned comparisons between the 2 methods may explain the difference between these two values. It is worthy to cite that our meta-analysis on 3DLASr is on a lower scale than previous meta-analyses on 2DLASr.Comparison of 2DLASr and 3DLASr in the two included studies revealed that 2DLASr was more than 3DLASr in both investigations, which is compatible with the findings of a previous meta-analysis14. Although it does not mean that factors such as age exert no effects on the LA reservoir function, it does indicate that these factors are not a source of variability in the normal reference value presented in the included studies. A reduction in 2DLASr in tandem with increasing age has been demonstrated in several studies31, but we stress that the age range in the studies included in the present meta-analysis was not wide (32–54 years). A previous investigation demonstrated an increase in 3DLASr after 50 years32. In most of the studies included in our meta-analysis, the participants were younger than 50 years of age30.We did not find age and sex as a source of heterogeneity, which chimes in with the finding of a previous meta-analysis regarding 2DLASr31. In a study that evaluated 3DLASr in men and women, the results showed no difference between the two sexes vis-à-vis this marker32. Likewise, we did not find sex to be a source of inter-study heterogeneity.Sex was not a determinant of 2DLASr in most studies subjected to meta-analysis30, which is concordant with our findings.Comparison of 2DLASr between European and Asian races in a previous investigation demonstrated no difference between the two ethnic groups14, only three studies reported the body surface area with a very narrow range in our meta-analysis.Whereas a meta-analysis on 2DLASr found a correlation between the body surface area and 2DLASr14. Nonetheless, several studies have demonstrated that the heart rate is not an independent determinant of 2DLASr31. Our results showed the heart rate as the source of inter-study heterogeneity. It is worthy of note that the range of the heart rate in the studies included in our meta-analysis was narrow (65–69 bpm), with only three studies having reported the heart rate as the mean and the standard deviation. It is, therefore, possible that the heart rate by chance presented itself as a source of inter-study heterogeneity.The heart rate is the source of heterogeneity in studies regarding 2DLASr31. The findings of our study are compatible with the studies reporting this association.Blood pressure in healthy subjects is associated with 2DLASr33. The LV ejection fraction is a marker of the LV systolic function, and 3DLASr is a marker of the LA reservoir function. In our study, the mean LV ejection fraction was within the normal range, so the LV ejection fraction was not a source of heterogeneity between the included studies. The LA volume index was correlated with 2DLASr in some studies31. The LA volume index in the included studies was lower than that in studies that confirmed the correlation between 2DLASr and the LA volume index31.The LV systolic function is correlated with the LA reservoir functionSubvolume numbers (4 vs 6) were not correlated with 3DLASr. The increased subvolume was accompanied by an increased frame rate as an index of temporal resolution, but it seems that it was not a determinant of heterogeneity between the included studies.Our exclusion of the single General Electric software-based study resulted in no significant changes in the normal range of LASr.We found the publication year to be a source of heterogeneity. It is possible that the time elapsed since the first 3DLASr measurement is allied to more expertise in this regard.The finding of the source of heterogeneity in our study by meta-regression was affected by the low number of studies included. In addition, not all the included studies provided the details of the study population such as the body mass index or the body surface area, clinical data such as blood pressure, or echocardiography data such as the LA volume index, precluding an in-depth analysis. The factors may, therefore, be a source of heterogeneity between studies.The results of our meta-analysis showed no significant publication bias, although the limited number of included studies precluded exact results. It should be, however, noted that we searched studies via a hierarchical method whereby three independent researchers reviewed all abstracts and selected full texts; additionally, all discrepancies were resolved according to consensus between two of the researchers. This checking method lessened the probability of missing related manuscripts.Our study results yielded the normal reference range of 3DLASr, although the range was obtained from a low number of suitable studies. Our findings can be used in clinical practice until future studies provide more robust data regarding 3DLASr among healthy normal subjects. These values may assist in the longitudinal assessment of patients who suffer from chronic conditions such as heart failure with reduced or preserved ejection fractions or those who were exposed to acute events such as myocardial infarction because they can depict changes from the normal range to the abnormal range.Our study revealed a need for more studies regarding the normal reference value of 3DLASr because of the dearth of high-quality studies in this field. Indeed, the majority of the studies that were suitable for our meta-analysis did not measure deformation in other directions or during the conduit or contraction phase. In addition, data are absent regarding the comparison of 3DLASr and the value obtained by cardiac magnetic resonance imaging.The present meta-analysis incorporated five studies on a total of 316 subjects. This sample size is low by comparison with studies via 2DSTE. We are hopeful that the publication of more studies investigating deformation in other directions and in other phasic times will contribute to more optimal meta-analyses since we were not able to provide values for these markers in our study.34.The studies included in our meta-analysis were observational or case–control, which are associated with high heterogeneity compared with randomized control trials because of the objective entity of these types of studiesThe software that is used for 3DSTE is semiautomatic, and the competency of the operators who obtain deformation data is unknown. In our meta-analysis, we considered the same value for all the studies included. The absence of large-scale studies yielding preliminary normal reference values should be addressed in the future.We could not conclusively assess the difference in vendors since we had four studies performed with Toshiba software and only one study conducted with General Electric software.The fact that the study populations in the case and control groups of all the studies incorporated in the current meta-analysis were Asian also prevented us from demonstrating any possible differences concerning the normal reference value of 3DLASr between ethnicities.35. Although this checklist prevents significant bias in the evaluation of studies, it is not objective totally36. What can also be deemed a limitation is that we did not exclude any studies because of low quality. Still, the overall quality of the included studies, albeit not excellent, was by no means poor insofar as they satisfied about 60% of the required items on average.Another weakness of note is that we analyzed study data, and not patient data, on account of the unavailability of the former. We also assessed the quality of the included studies according to the checklist used in similar meta-analyses in this field37.Finally, we were not able to conduct a stratified meta-analysis by considering the quality of studies because of the innate properties of qualityWe demonstrated a mean global 3DSTE-derived longitudinal LASr of 27.5%  in normal healthy subjects. The heterogeneity in different normal reference values was related to the publication year, the heart rate, and systolic and diastolic blood pressure. Further studies in this field are required to confirm the normal reference value of the LA deformation markers obtained by 3DSTE.Supplementary Information.Supplementary Table S1.Supplementary Table S2.


Bayesian methods are an important set of tools for performing meta-analyses. They avoid some potentially unrealistic assumptions that are required by conventional frequentist methods. More importantly, meta-analysts can incorporate prior information from many sources, including experts’ opinions and prior meta-analyses. Nevertheless, Bayesian methods are used less frequently than conventional frequentist methods, primarily because of the need for nontrivial statistical coding, while frequentist approaches can be implemented via many user-friendly software packages. This article aims at providing a practical review of implementations for Bayesian meta-analyses with various prior distributions. We present Bayesian methods for meta-analyses with the focus on odds ratio for binary outcomes. We summarize various commonly used prior distribution choices for the between-studies heterogeneity variance, a critical parameter in meta-analyses. They include the inverse-gamma, uniform, and half-normal distributions, as well as evidence-based informative log-normal priors. Five real-world examples are presented to illustrate their performance. We provide all of the statistical code for future use by practitioners. Under certain circumstances, Bayesian methods can produce markedly different results from those by frequentist methods, including a change in decision on statistical significance. When data information is limited, the choice of priors may have a large impact on meta-analytic results, in which case sensitivity analyses are recommended. Moreover, the algorithm for implementing Bayesian analyses may not converge for extremely sparse data; caution is needed in interpreting respective results. As such, convergence should be routinely examined. When select statistical assumptions that are made by conventional frequentist methods are violated, Bayesian methods provide a reliable alternative to perform a meta-analysis. Systematic reviews and meta-analyses have been a popular tool for synthesizing evidence in many fields, including evidence-based medicine, public health, and environmental research ,2,3. ThrComing along with the mass production of meta-analyses in recent years, the quality of many meta-analyses may be low, being partly related to certain methodological limitations . For exaoutcomes ,11,12,13outcomes ,15,16,17When comparing frequentist approaches to meta-analysis, Bayesian methods are widely used in medical research among other fields, and they provide a flexible way for implementing complicated models ,22,23,24The BMJ. Based on these examples, we empirically examine the impact of different priors on the overall effect estimates, including point and interval estimates, and explore potential problems that are caused by sparse data.The objectives of this article are three-fold, with a focus on univariate meta-analyses of ORs for binary outcomes. First, we review Bayesian methods, as well as frequentist methods, for conducting meta-analyses. As the heterogeneity of effect sizes plays a critical role when combining primary study results, we introduce three commonly used estimators of the between-studies variance under the frequentist framework. Under a Bayesian framework, we present four prior distributions with different sets of hyper-parameters. It is of note that this manuscript does not aim to provide recommendations for prior selection, because preferable priors may differ case-by-case, particularly when experts’ opinions are available to derive informative priors. Second, we present detailed steps to implement a Bayesian meta-analysis via the Markov chain Monte Carlo (MCMC), with an emphasis on the importance of validating the posterior samples . Third, we apply the various methods to data from five meta-analyses that were published in The In order to implement Bayesian random-effects meta-analysis of ORs, we consider a meta-analysis containing model is ,42,43,44d trials ,45,46,47Using the logit link function The vague normal prior e for τ2 , choosin) length .We consider four types of priors for the heterogeneity parameter: inverse-gamma, uniform, half-normal, and log-normal priors. Different hyper-parameters are used for each prior distribution. We choose these priors, because they have been used in many Bayesian meta-analyses. One may also tailor a specific prior using experts’ opinions on a case-by-case basis; such practice is out of the scope of this article.The inverse-gamma prior, which is denoted by putation . Small ppractice . The invpractice . NeverthThe uniform prior, denoted by iation τ , where c and 100 . The jusThe half-normal prior, which is denoted by or for τ ,53. This 1, or 2 .In addition, the evidence-based informative log-normal prior has been suggested for τ2 . We denoVarious approaches, including trace and density plots, are available for diagnosing MCMC samples and their possible convergence. A trace plot presents the posterior samples after the burn-in period in each Markov chain iteration for each parameter. If it shows well-mixed samples, this provides evidence that the Markov chains have stabilized and converged. A density plot depicts the posterior sample density for each parameter. If the density shows multiple modes , then the Markov chains may not have converged. When Markov chains’ convergence and stabilization have been justified, point estimates , and their 95% CrIs, often formed by 2.5% and 97.5% posterior quantiles, can be computed for the Bayesian meta-analysis.We used five real-world datasets with different sizes (number of studies and within-study sample sizes), event rates, and heterogeneity extents to illustrate the implementation of Bayesian meta-analyses, the impact of prior distributions on meta-analytic results, and the potential problems of MCMC convergence. All of the datasets had binary outcomes; regardless of original analyses, (log) ORs were used as effect measures in our re-analyses.Example 1. The meta-analysis by Lamont et al.  investigExample 2. The meta-analysis by Crocker et al.  combinedExample 3. The meta-analysis, as reported by Baxi et al. , includeExample 4. This meta-analysis is also from Baxi et al. ; it combExample 5. The fifth meta-analysis by Martineau et al.  collecteFor each dataset, we implemented the random-effects meta-analysis using the DerSimonian–Laird (DL) , maximumIn the presence of zero event counts, when implementing the conventional frequentist methods, studies with zero counts in both the treatment and control groups had to be excluded (see We used the R package “rjags” (version 4–7) to implement the Bayesian models via the MCMC algorithm. For each Bayesian model, we used three Markov chains, each having a burn-in period of 50,000 iterations, followed by 200,000 iterations for drawing posterior samples with thinning rate 2. Trace plots were used to assess the chains’ convergence. We obtained the posterior medians of the overall OR, the heterogeneity standard deviation metafor” . We useds) of τ^ . The SupThis study did not require ethical approval and patient consent, because it focused on statistical methods for meta-analyses, and all the analyses were performed using publicly available data.The estimated overall ORs by all methods ranged from 4.30 to 4.59, and their CIs/CrIs were all above 1, which suggested a significantly increased risk of recurrent stillbirth among women with a previous stillbirth compared with those with a previous live birth. The results were relatively robust to the different meta-analytic methods, likely because this meta-analysis contained a moderate number of studies , the choice of prior may have a noteworthy impact on meta-analysis results. Ideally, meta-analyses may use priors that represent experts’ opinions. If such priors are unavailable or disputed, sensitivity analyses that use different candidate priors are recommended. In addition, meta-analysts should routinely examine the convergence and stabilization of MCMC processes when performing Bayesian meta-analyses; this is often overlooked in current practice, as illustrated in Examples 3 and 4. When event rates are low and zero event counts appear in many arms in a meta-analysis, the MCMC algorithm may not converge; as such, the inference that is based on posterior samples is unreliable. We have summarized several possible solutions to handle this problem in Bayesian methods for meta-analysis are not problem-free, and their results must be carefully validated before being applied to decision-making, as discussed above. This article did not intend to promote Bayesian meta-analyses to all situations and involve in the debates on frequentist vs. Bayesian inference. Instead, we aimed at providing some practical guidelines for researchers who are interested in implementing Bayesian meta-analyses. One may refer to other papers that especially aimed at comparing these two types of methods to better understand the pros and cons of frequentist and Bayesian meta-analyses ,82,83,84This article limited the review of Bayesian methods to the case of ORs. Bayesian methods are broadly available for meta-analyses of other effect measures  with binary outcomes and for meta-analyses with other outcomes  ,85. Simi


The role of lipoprotein-A [Lp (a)] as a risk factor for stroke is less well documented than for coronary heart disease. Hence, we conducted a systematic review and meta-analysis for the published observational studies in order to investigate the association of Lp (a) levels with the risk of stroke and its subtypes. In our meta-analysis, 41 studies involving 7874 ischemic stroke (IS) patients and 32,138 controls; 13 studies for the IS subtypes based on TOAST classification and 7 studies with 871 Intracerebral hemorrhage (ICH) cases and 2865 control subjects were included. A significant association between increased levels of Lp (a) and risk of IS as compared to control subjects was observed (standardized mean difference (SMD) 0.76; 95% confidence interval (CIs) 0.53–0.99). Lp (a) levels were also found to be significantly associated with the risk of large artery atherosclerosis (LAA) subtype of IS  as well as significantly associated with the risk of ICH  as compared to controls. Increased Lp (a) levels could be considered as a predictive marker for identifying individuals who are at risk of developing IS, LAA and ICH. Almost 80% of strokes are ischemic stroke (IS) and 15–20% are haemorrhagic stroke (HS) in origin3. According to the Trial of Org 10172 in Acute Stroke Treatment (TOAST) classification; IS has been categorised according to the presumed etiological mechanism into five groups: large artery atherosclerosis (LAA), small vessel disease (SVD), cardioembolic disease (CE), other determined etiology (ODE), and undetermined etiology (UDE)4.Stroke is reported as the most common cause of long term disability and the second most leading cause of death worldwide6. By binding LDL, calcium, and other components into an atherosclerotic plaque on the walls of blood arteries, Lp (a) is hypothesised to speed up the development of atherosclerosis7. The LPA gene regulates the variation in Lp (a) plasma concentrations genetically, ranging from 36% in the PROCARDIS8 consortium to 70–90% in genome-wide association studies, with larger apo(a) isoforms related with lower values of Lp (a)10. The concentrations of Lp (a) range from 0.1 mg/dl to more than 200 mg/dl11.Lipoprotein (a) or Lp (a) is a lipoprotein moiety that consists of core lipoprotein molecule, containing apolipoprotein B (apo-B100), to which a glycoprotein of variable molecular weight, apolipoprotein (a) [apo(a)], is covalently attached via a cysteine-cysteine disulfide bond12. Plasma Lp (a) concentrations appear to be regulated by synthesis rather than catabolism, according to kinetic studies. Concentration and pathological responses may be influenced by apo(a) sequence polymorphisms. Lp (a)/apo(a) functions may also be affected by changes in circulating Lp (a). Importantly, the relevance of apo(a) in cardiovascular diseases (CVD s) and peripheral vascular disorders, as well as its physiological function, remain unknown, and there is no effective therapeutic option for decreasing increased Lp levels (a)15.On a cellular level, apo(a) undergoes post-translational changes in the endoplasmic reticulum as a secretory protein (ER). The length of time it takes to modify larger apo(a) isoforms is determined by the size of the apo(a) isoform. As a result, larger apo(a) molecules are produced at a slower rate per unit of time, resulting in lower Lp (a) plasma concentrations19, while others have found no link22. This could be due to a lack of discrimination across incidence stroke subtypes22, as well as ethnic or other disparities in cohort composition. A growing number of epidemiological studies have found a link between dyslipidemia and atherosclerosis-related stroke. Indeed, the lipid metabolism of different stroke types and IS subtypes differs dramatically26. Two previously published meta-analyses28 had confirmed that elevated Lp (a) is an independent risk factor for IS, however, IS subtypes based on TOAST classification as well as HS remains to be explored further. Hence, we conducted a systematic review and meta-analysis for the published observational studies in order to investigate the association of Lp (a) levels with the risk of stroke and its subtypes.Other large, population-based cohort studies on stroke have produced mixed results, with some research associating raised Lp (a) levels to a higher incidence of IS29. A comprehensive search for all the published articles was performed in electronic databases including PubMed, EMBASE, Cochrane Library, Trip Databases, Worldwide Science, and Google Scholar from 01st January 1950 to 30th April 2020. Following search terms: ‘Lipoprotein (a)’ OR ‘Lp (a) Levels’ OR ‘Lipid Biomarkers’ AND ‘Stroke’ OR ‘Subtypes’ OR ‘TOAST Classification’ OR ‘Ischemic Stroke’ AND ‘Haemorrhagic Stroke’ OR Intracerebral Haemorrhage’ OR ‘ICH’ AND ‘Cerebrovascular Disease’ AND ‘cerebral infarction’ were used. Reference lists of the selected studies were also searched manually to obtain any additional eligible studies on human subjects. No restrictions related to language, sex and publication year was applied.This systematic literature review was performed using the guidelines of the PRISMA statement (1) Observational studies including case–control, nested case–control, cross sectional and cohort design investigating the association of Lp (a) levels with the risk of stroke or stroke types or IS sub-types based on TOAST classification compared to control subjects; (2) studies with clinically confirmed diagnosis of stroke (ischemic or haemorrhagic) using CT or MRI scans; (3) patients aged ≥ 18 years (adult population); (4) studies reporting numbers for patients and control groups as well as raw values for Lp (a) levels.(1) Duplicates, case reports, case series, systematic reviews, conference abstracts, preprints and editorials; (2) Studies not reporting relevant outcomes; (3) Unavailability of full-texts.30. The assessment criteria involving NOS uses three broad criteria viz. selection, comparability and exposure. Selection criteria defines and analyses the cases and control subjects included in the study, comparability defines the matching or comparison of cases and control subjects for better empirical investigation and exposure determines whether the study was conducted in a blinded or unbiased manner along with the response of the subjects. Publication bias was assessed using Begg’s and Egger’s funnel plot analysis32.The risk of bias was assessed by Newcastle–Ottawa Scale (NOS) for quality assessment of all the included studies in the meta-analysishttp://unitslab.com/node/85).All relevant studies were analysed separately by two reviewers (PK and PS) based on the inclusion criteria listed above. The analysis was done first at the title and abstract level and then at the full-text level. Any disagreement was resolved by discussion with a third reviewer. Following data were extracted from the studies which included: First Author’s Name, Published Year, Ethnicity, Country, Study Design, Number of Cases and Controls, Mean Age, mean and standard deviation values of biochemical parameters including Lp (a), methods of Lp (a) assay and follow up duration. Data was extracted independently by two authors (PK and PS) using a standardized extraction table. Lp (a) concentrations were reported with different units in the included studies and were converted to similar units for analysis purpose using online unit conversion tools .A random or fixed effect model was used to calculate the pooled Standardized Mean Difference (SMD) or Odds Ratio (OR) with 95% confidence interval (CI). Heterogeneity was calculated with the Ethical approval was not required for this manuscript as it was a systematic review and meta-analysis done by using the existing published data and the research was not directly conducted in any human subjects.Figure 71. Thirty-five studies were case–control, one was nested case–control and five were population-based cohort studies. The publication years of the studies included in our meta-analysis ranged from 1985 to 2020. The studies were divided into two groups of populations based on ethnicity; 25 studies were conducted in Caucasian population and 16 studies were in Asian population. The sample size for IS cases ranged from 31 to 1953. Twenty-two studies used hospital-based (HB) source of control and nineteen studies used population-based (PB) source of control. The quality score was high in nine articles, medium in 22 articles and low in 10 articles. The detailed quality scores (NOS scores) of the included studies ranging from low to high are represented in supplementary tables  with the risk of IS as compared to control subjects with a total 7874 IS cases and 32,138 control subjects55 reported calculated OR with 95% CI for the association of Lp (a) with the risk of IS as compared to control subjects; which have been used directly for estimating pooled ORs with 95% CI.ELISA technique was found to be most common methods for LPA assay and reported in 18 studies, follow-up duration and LPA timepoints were reported in limited studies. No information was available for the stroke patients undergoing any LPA treatments in all the included studies in our meta-analysis. Only 11 studies75 reported the association of Lp (a) with the risk of IS subtypes based on TOAST classification. The publication years ranged from 1992 to 2019. Out of 13 studies, eight studies reported data for LAA subtypes74; nine studies for SVD subtypes74 and five studies for CE subtypes66 with control subjects. Seven studies were conducted in Caucasian population and six studies were in Asian population. The sample size for IS subtypes ranged from 09 to 1809. Ten studies were case–control, one was cross-sectional and two were population-based cohort studies. The quality score was medium in nine studies, high in two studies and low in two studies as represented in supplementary tables  cases and 2865 control subjects were identified for the association of Lp (a) levels with the risk of ICH as compared to control subjectsA significant association between increased levels of Lp (a) and risk of IS as compared to control subjects was observed . A subgroup analysis based on ethnicity also showed a significant association between increased levels of Lp (a) and the risk of IS in 16 Asian  as well as 25 Caucasian studies  respectively  and the risk of IS as compared to control groups in 35 case–control studies  and one nested case–control study   and the risk of IS as compared to control groups in high-quality studies ; medium  and low-quality studies   with the risk of IS as compared to control groups . Based on ethnicity, a significant association of increased levels of Lp (a) with the risk of IS as compared to control groups was observed for Asian population  but not for Caucasian population . A significant association of increased levels of Lp (a) with the risk of IS as compared to control groups  was observed for case–control studies but not for prospective cohort studies . High-quality studies confirmed a significant association of increased levels of Lp (a) with the risk of IS as compared to control groups  but not the medium  and low-quality studies   levels with the risk of LAA stroke subtype vs. control was investigated in eight studies and our findings reveal an overall significant association of increased levels of Lp (a) with the risk of LAA as compared to control groups . Based on ethnicity, no significant association for the increased levels of Lp (a) with the risk of LAA in Asian population  as well as Caucasian population  was observed. All included eight studies were of case–control design which showed a significant association and based on NOS quality grading, no association was observed for medium quality studies . Single studies were found based on high and low NOS quality and reported significant association for the increased levels of Lp (a) with the risk of LAA as compared to control groups (Table No significant association for the increased Lp (a) levels with the risk of SVD subtype vs. control  was observed. Moreover, based on ethnicity, study design and NOS quality grading, similar non-significant association was observed except for high quality study which included only a single study for SVD vs. control subjects  (Table No significant association for the increased Lp (a) levels with the risk of CE stroke of IS subtype vs. control  was observed. Subgroup analysis based on ethnicity, study design and NOS quality grading also revealed a non-significant association for the increased Lp (a) levels with the risk of CE subtype vs. control subjects  levels with the risk of ICH as compared to control subjects  was observed Fig. . Non-sigThe shape of the funnel plots indicated the presence of publication bias while analysing the Lp (a) levels with the overall risk of IS. After conducting the Begg’s test, we observed that a significant publication bias was present in the included studies for Lp (a) levels with the overall risk of IS  Figure a. The shTo further explore the amount of heterogeneity present in our meta-analysis, we conducted meta-regression analysis based on NOS quality score, study design and ethnicity for determining the impact of heterogeneity. Significant heterogeneity was observed for overall IS vs. control based on NOS quality (p = 0.005)  investigating for the association of Lp (a) with LAA vs. control; two studies  for SVD vs. control; three studies  for CE vs. control; three studies  for ICH vs. Control which could have potentially affected the overall effect size estimates  levels and its association with the risk of IS, IS subtypes based on TOAST classification and ICH compared to control subjects. To the best of our knowledge, this is the most robust and the largest meta-analysis conducted till date which comprised of both Asian and Caucasian ethnicities to ascertain the risk of IS, IS subtypes and HS with increased levels of Lp (a). Our meta-analysis revealed that increased levels of Lp (a) are significantly associated with the risk of IS in Asian as well as Caucasian population. Also a significant association for the increased level of Lp (a) with the risk of LAA and ICH as compared to control subjects was observed. Lp (a) levels were found to be greater in Asian population as compared to Caucasian population confirming greater risk of stroke as compared to control group. However, on the basis of subtypes, no significant association was observed in Asian as well as Caucasian population separately for increased levels of Lp (a) in LAA, SVD and CE subtypes when compared to control groups.28. A total of thirty-one studies were included in the meta-analysis by Smolders et al. 200727 in which the association of overall stroke was found to be statistically significant with Lp (a) increment levels  which is in agreement with the findings of our meta-analysis. Another meta-analysis by Nave et al. 201528 observed a significant association between Lp (a) and IS with OR of 1.41 (95% CI 1.26–1.57) for case–control studies (n = 11) and the pooled estimated risk ratio was 1.29 (95% CI 1.06–1.58) for prospective studies (n = 9).Two previous meta-analyses had also established the association of elevated Lp (a) levels and risk of stroke by pooling data from case–control, prospective cohort and nested case–control studiesDespite the fact that this systematic review and meta-analysis was undertaken comprehensively with defined inclusion and exclusion criteria along with uniform measured-effect across all analyses, the study has some following limitations: 1)included studies had a wide range of incorporated variables like age, ethnicity, sample size, study-design; 2) mean and standard deviations of Lp (a) levels obtained from few studies were converted from either the actual reported median values or the inter-quartile range values, inferring that they did not actually represent the original mean and standard deviation values of Lp (a) levels. (c) Subgroup analysis based on cut-off values of Lp (a) levels was not performed owing to non-availability of cut-off values of Lp (a) in majority of the included articles as represented in Tables included  mean andIncreased Lp (a) levels could be considered as a predictive marker for identifying individuals who are at risk of developing IS, LAA and ICH.Supplementary Tables.Supplementary Figures.


Motivated by the increasing incidences of skin cancer, in 2015, Australian states banned indoor tanning to prevent exposure to artificial ultraviolet light. However, there has been no study investigating the association between indoor tanning and early-onset melanoma and non-melanoma skin cancer. In this study, we reviewed a total of 54 studies to examine the association between indoor tanning device use and overall and early-onset skin cancer. We found that indoor tanning is associated with increased risk for early-onset melanoma and NMSC, and has a dose–response relationship with first exposure at an early age and the frequency of exposure. Therefore, this study emphasizes the importance of avoiding indoor tanning risk in younger adults. Our findings provide evidence that supports policies regulating the excessive use of tanning devices, especially in the vulnerable younger population, to reduce the additional risk of skin cancer.The aim of this study was to examine the association between indoor tanning use and the risk of overall and early-onset (age < 50) melanoma and non-melanoma skin cancer (NMSC). To evaluate the association between indoor tanning and skin cancer, a systematic review of the literature published until July 2021 was performed using PubMed, EMBASE, and MEDLINE. Summary relative risk (RR) from 18 studies with 10,406 NMSC cases and 36 studies with 14,583 melanoma cases showed significant association between skin cancer and indoor tanning , RR = 1.58, 95% CI 1.38–1.81; basal cell carcinoma (BCC), RR = 1.24, 95% CI 1.00–1.55). The risk was more pronounced in early-onset skin cancer . Moreover, first exposure at an early age (age ≤ 20 years) and higher exposure  to indoor tanning showed increasing risk for melanoma  and NMSC . These findings provide evidence supporting primary prevention policies regulating modifiable behaviors to reduce the additional risk of skin cancer among younger adults. Skin cancer, including melanoma and non-melanoma skin cancer (NMSC), is the fifth most commonly occurring cancer, with over 1 million diagnoses worldwide and incidence rates continuing to increase rapidly . The shaExposure to ultraviolet (UV) radiation is a major environmental factor for melanoma and NMSC. According to previous studies, almost 90% of all melanomas, 85% of squamous cell carcinoma (SCC), and 82% of basal cell carcinoma (BCC) were attributable to excess UV radiation ,8. AccorThis systematic review aims to assess and delineate the indoor tanning device type, as well as the clinicopathological characteristics of indoor tanning-associated overall and early-onset melanoma and NMSC. This study searched publications published up to July 2021 using the databases PubMed, EMBASE, and MEDLINE. The inclusion criteria for eligibility were as follows: (1) for the outcome of interest, the following keywords were considered: “skin cancer”, “skin neoplasm”, “melanoma”, “squamous cell carcinoma”, “SCC”, “basal cell carcinoma”, and “BCC”; (2) for exposure, following keywords were used: “sunbed”, “sunlamp”, “sunbathing”, “tanning”, “UV”, “artificial UV”, “artificial light”, “solaria”, and “solarium”; and (3) providing data for the calculation of the relative risk (RR) or odds ratio (OR) with 95% confidence intervals (CIs). All the titles and abstracts of the searched papers were reviewed to identify potentially eligible studies, and, furthermore, all the references from previous meta-analyses on skin cancer written in English were carried out. Cohort and case–control studies published as original articles were also selected in this study. For cohort studies from the same population, an updated dataset was included in this study. All included studies examined patients given the diagnosis of NMSC or melanoma who were ever exposed to an indoor tanning device. This study was conducted by following the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) . This stThe pooled relative risk (RR) and 95% confidence intervals (95% CIs) of each association were calculated based on the risk with the whole confounding adjustment. When no estimates were suggested, the calculated crude estimates with 95% CIs obtained from the reported data were included in this study. For studies with estimates for both SCC and BCC, combined RRs of SCC risk and BCC risk were calculated as RRs of NMSC.2 statistics. Funnel plots and Egger and Begg tests were used to assess potential publication bias derived from the validity of the estimates. All analyses were done with R version 4.0.2. The meta-analysis used the random effects models using the maximum likelihood estimation to estimate the summary RR. Statistical heterogeneity evaluated the heterogeneity of the studies using Higgins and Thompson’s IThe following characteristics were analyzed from each study: (1) onset age ), (2) histology , nodular melanoma (NM), lentigo maligna melanoma (LM), and others for melanoma); (3) anatomic site ; (4) type of indoor tanning device (sunbed vs. sunlamp); (5) study design (cohort vs. case–control study); (6) publication year (<2000 vs. ≥2000); (7) first exposure at early age to indoor tanning (<20 years vs. ≥20 years); (9) annual frequency of indoor tanning (<10 times vs. ≥10 times). According to the research strategy, a total of 484 records were identified. Of those articles, 399 were excluded by titles and abstracts, and 85 full-text articles were reviewed to assess their eligibility. Twelve studies with different exposures and outcomes, ten studies that were not original articles, five studies with duplicated populations, and four descriptive studies were removed from the 85 full-text articles. Finally, a total of 54 eligible articles  were selected for the meta-analysis. More details of publication selection for this meta-analysis are shown in A total of 10,406 NMSC cases from 18 studies were included in this study. The summary RRs of indoor tanning device use were 1.40 (95% CI 1.18–1.65) for overall NMSC, 1.58 (95% CI 1.38–1.81) for SCC, and 1.24 (95% CI 1.00–1.55) for BCC. Based on the 36 studies including 14,583 melanoma cases, the summary RR of indoor tanning device use was 1.27 (95% CI 1.16–1.39) for overall cutaneous melanoma. These associations were even greater for early-onset skin cancer, defined as melanoma and NMSC occurring under the age of 50. The summary RR for early-onset NMSC was 1.81 (95% CI 1.38–2.37), for SCC was 1.99 (95% CI 1.48–2.68), and for BCC was 1.79, (95% CI 1.14–2.69). The summary RR for early-onset melanoma was 1.75 (95% CI 1.14–2.69) compared with never-users . According to the type of indoor tanning device, both sunlamp  and sunbed use  were significantly associated with the increasing risk of SCC. For BCC, a significant association was observed in sunbed exposure , while no significant association was shown in sunlamp use  . After sThe anatomic sites of melanoma were classified by trunk, head and neck, and limbs. After stratifying the anatomic sites, ever-use of indoor tanning increased the risk of melanoma on the trunk  and on the limbs  . For his2 = 7%; p = 0.37) and 1.30 (95% CI 1.14–1.47) for 31 case–control studies. There was also a consistently significant association in before  and after 2000   and sunbed exposure  were significantly associated with the risk of melanoma . For an 12–1.42) . To assess the dose–response of indoor tanning, sub-group analyses were conducted, including first exposure at an early age to indoor tanning and the annual frequency of indoor tanning use. This study found that first exposure at an early age to indoor tanning was associated with 1.47-fold higher risk for melanoma (95% CI 1.16–1.85). High dose exposure to indoor tanning more than 10 times per year also related to an increasing risk of melanoma . For NMSC, the RR of first exposure at an early age was 2.02 (95% CI 1.44–2.83) for NMSC, 1.89 (95% CI 0.90–3.98) for SCC, and 1.86 (95% CI 1.44–2.41) for BCC. Compared to never-users, more than 10 tanning times a year had significantly increased the risk of NMSC , SCC , and BCC  . This meta-analysis provides evidence of the association between indoor tanning and melanoma and non-melanoma skin cancer, particularly early-onset skin cancer diagnosed before age 50 years. These associations were stronger for melanoma occurring on the trunk and limbs. By stratifying the types of indoor tanning, this study found that both sunlamps and sunbeds were significantly associated with the risk of melanoma and squamous cell carcinoma. Moreover, first exposure at an early-age to indoor tanning, and frequent use of indoor tanning was significantly associated with an increased risk for melanoma and non-melanoma skin cancer. Although there are several previous meta-analyses of the association between indoor tanning device use and skin cancer, the association was mostly based on population-based studies ,17,18,19In 2006, the IARC evaluated the dose–response relation between sunbed use and the risk of melanoma . AlthougThe type of indoor tanning device is divided into sunbed and sunlamp. It is controversial that sunlamps and sunbeds have different effects on skin cancer. A population-based study showed modest association between sunlamp use and melanoma risk, whereas no association was found for sunbeds . On the Melanoma and NMSC mainly have different anatomical locations where cancer occurs. Melanoma is commonly diagnosed in the trunk ,35,36, wBesides, this meta-analysis confirmed the consistency that individuals who had first exposure at an early age to indoor tanning (<20 years) and had more than 10 tanning times in a year were associated with skin cancer risk. This is in line with the previous research that showed a dose–response relation between indoor tanning and the risk of melanoma and NMSC ,42,43,44Our study has several limitations. First, individual studies included in this meta-analysis were concentrated in the population from Europe and North America. Second, due to the limited information of histological subtypes of melanoma in prior studies, the risk of histological melanoma subtypes from indoor tanning was observed as mostly insignificant in our meta-analysis. Third, the insufficient number of SCC and BCC studies with information on the tanning type. Although we found a significant association between sunbed usage and SCC and BCC, there may not be enough statistical power to verify the link. Further meta-analysis should be performed based on more studies in the future to verify this association for SCC and BCC. Nevertheless, this meta-analysis has several strengths. Unlike previous studies, this study presents the risk of early-onset and histological and anatomical subtypes by exposure to indoor tanning according to device type. In conclusion, this meta-analysis suggests indoor tanning increases the risk of skin cancer, particularly early-onset melanoma and NMSC. There are constantly increasing trends for skin cancer worldwide , especia


Although a considerable volume of data supporting induction or aggravation of psoriasis because of angiotensin-converting enzyme (ACE) inhibitor use exists, it remains insufficient for definitive conclusions. Therefore, we aimed to evaluate the association between ACE inhibitor use and psoriasis incidence through a systematic literature review and meta-analysis. We searched for qualifying studies across PubMed, Web of Science, and Embase. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to evaluate the strength of the association between ACE inhibitor use and psoriasis incidence. Eight studies with a total of 54,509 patients with a psoriasis diagnosis were included in this meta-analysis. The pooled OR for psoriasis incidence among ACE inhibitor users was 1.52  compared to that among non-users. From subgroup analysis by continent, the OR for ACE inhibitor users versus non-users was 2.37 (95% CI 1.28–4.37) in Asia. Per the subgroup analysis by climate, the OR for ACE inhibitor users vs non-users in dry climate was 3.45 (95% CI: 2.05–5.79) vs 1.32 (95% CI 1.01–1.73) in temperate climate. Our results reveal a significant association between ACE inhibitor use and psoriasis incidence. Although ACE inhibitors are generally well-tolerated, several adverse events have been reported, including dry cough, angioedema, rash, and hyperkalemia2. In addition, psoriasis has been reported as a rare adverse event. Several case reports have implicated the use ACE inhibitors, particularly captopril and enalapril, in the incidence and exacerbation of psoriasis9. Moreover, psoriasiform eruptions and exacerbation of pre-existing psoriasis were reported in patients receiving other ACE inhibitors such as ramipril and lisinopril5.Angiotensin-converting enzyme (ACE) inhibitors are widely used medications for treatment of hypertension and congestive heart failure and prevention of kidney diseases in patients with hypertension or diabetes10.Therapeutic agents which may induce or aggravate psoriasis may be classified into three categories depending on the strength of the association. First group of drugs have a strong association with psoriasis incidence or aggravation and include lithium, β-blockers, and antimalarials. Second group of drugs have considerable yet insufficient data supporting incidence or aggravation of psoriasis, such as ACE inhibitors, interferons, and terbinafine. Third group of drugs have been occasionally reported to be associated with incidence or aggravation of psoriasis and include a large group of miscellaneous drugs such as clonidine, digoxin, and amiodaroneAlthough studies reporting the possible association between ACE inhibitor use and psoriasis exist, the results have been inconsistent. Therefore, we investigated the association between ACE inhibitor use and psoriasis incidence through a systematic literature review and meta-analysis. In addition, because environmental factors may affect the association between ACE inhibitor use and psoriasis, subgroup analyses by study design, continent and climate were also performed.18.The study selection process is summarized in Fig. 18 and four were case–control studies14. Four studies were performed in Asia18, three in Europe16, and one in the USA15. The number of patients in each study ranged from 625 to 256,356. Regarding study quality using the Newcastle–Ottawa Scale (NOS) score, each study received ≥ 6 points out of 9 ; hence, the random-effects model was used to calculate effect size. As there were four studies each in case–control and cohort study, we performed a subgroup analysis by study design. The ORs were 2.20 (95% CI 1.08–4.48) for case–control studies and 1.24 (95% CI 0.87–1.75) for cohort studies.For the association of ACE inhibitor use with psoriasis incidence, eight studies with a total of 54,509 patients with a psoriasis diagnosis were included in the meta-analysis. As shown in Fig. 18 and Europe16, respectively. As shown in Fig. For a subgroup analysis by continent, we divided into two groups; four and three studies in Asia18 and dry climates14, respectively. As shown in Fig. p = 0.05).For the subgroup analysis by climate, since there were five and two were conducted in temperate climatep = 0.50) nor the Begg’s test  showed significant publication bias for these studies. Sensitivity analysis was performed by sequentially excluding each study. There were no significant effects on ORs for the association of ACE inhibitor use with psoriasis and heterogeneity .Neither the Egger’s test , and epoxyeicosatrienoic acids. In contrast, inhibition of bradykinin degradation by ACE inhibitors alters the kinin-kallikrein arachidonic acid system leading to increased concentrations of inflammatory metabolites. This kinin-kallikrein arachidonic acid system is an important pathway in the pathogenesis of psoriasis23.A hypothesis for the association between ACE inhibitor use and psoriasis is related to bradykinin. ACE has two natural substrates and two catalytic domains: cleaving angiotensin I and inactivating bradykinin24. Moreover, angiotensin II stimulates the production of specific cytokines such as interleukin-12, potent monocyte chemoattractant protein-1, NO, and tumor necrosis factor-α, thereby causing infiltration of inflammatory cells into tissues25. A previous in vivo study observed that angiotensin II enhances vascular inflammation by activating nuclear factor-kappa B (NF-κB)-mediated pro-inflammatory genes. Angiotensin II also downregulates peroxisome proliferator-activated receptors (PPARs) alpha and gamma, resulting in decrease of the anti-inflammatory potential of PPARs26.Angiotensin II is known to have a proinflammatory activity, which is mediated by the activation of macrophages, T-cells, mesangial cells, dendritic cells, and vascular smooth muscle cells27. Binding of angiotensin II to AT1R increases inflammatory response whereas binding to AT2R decreases it. A recent study demonstrated that AT2R stimulation reduced expression of proinflammatory cytokine genes and activation of NF-κ in rheumatoid synovitis28. The use of ACE inhibitors prevents angiotensin II from binding to both receptors. Therefore, psoriasis due to ACE inhibitor use could involve more complex processes. A further study of this pathway is needed.The increased risk of psoriasis by the use of ACE inhibitors, although angiotensin II has aforementioned proinflammatory properties, is partially explained by the various activities of angiotensin II receptors. Angiotensin II has two major G protein-coupled receptor subtypes, the angiotensin II type 1 receptor (AT1R) and the angiotensin II type 2 receptor (AT2R)29. In a meta-analysis, the incidence of cough due to ACE inhibitor use has been reported to be 2.7-fold higher in East Asian populations compared to that in Caucasian populations30. These findings suggest the possible association between race and ACE inhibitor-related cough. Because bradykinin is seemingly the main mechanism of both ACE inhibitor-induced psoriasis and cough, the higher incidence of ACE inhibitor-related cough in Asians could explain the high risk of psoriasis in Asians.A previous study showed that Chinese patients had a notably higher risk of ACE inhibitor-related cough than Caucasians31. In general, psoriasis is more common in colder northern regions than in tropical regions. Incidence among African blacks from the hot, wet west-central countries is less than that among African blacks from the milder, dry east-central nations31. Consistent with this finding, the climate subgroup analysis in our study showed a stronger association between ACE inhibitor use and psoriasis in dry climates.Psoriasis is a multifactorial disease with both genetic and environmental contributing factors to its clinical manifestationThis meta-analysis has some limitations that should be considered when interpreting the results. First, considerable heterogeneity was observed. Second, some confounding factors, which could affect the risk of psoriasis, could not be adjusted for. For example, most patients using ACE inhibitors for hypertension or other indications generally have more than one commitment medications, which could affect study results. Besides, within-study variation remained even after subgroup analysis due to the lack of adjusted odds ratio data in the included studies. Third, the duration and dosage of the drug were not specified. Fourth, the power of Begg's and Egger's test was low due to the number of included studies was small (less than 10). Therefore, the results of Begg's and Egger's test need to be interpreted with caution. Although we conducted subgroup analyses to look into the between-study variations, they still have within-study variations.Nevertheless, this study is valuable in that it is the first systematic review and meta-analysis to evaluate the association between ACE inhibitor use and psoriasis. By combining the statistically insignificant findings of several studies, this study revealed significant associations between ACE inhibitor use and psoriasis. More prominent association between ACEI inhibitor use and psoriasis incidence was found in studies conducted in Asia or dry climate area. Based on our results, ACE inhibitor users should be carefully screened for cutaneous adverse events, particularly psoriasis.The literature search included querying PubMed, Embase, and Web of Science for studies up to August 6, 2020, on the association between ACE inhibitor use and psoriasis incidence. The search included the following keywords: ((psoriasis OR psoria* OR psoriatic) AND ((ACEI OR ACE inhibitor* OR angiotensin converting enzyme inhibitor*) OR (antihypertensive agent* OR antihypertensive drug* OR antihypertensive medication*)). After removing duplicates, two researchers independently screened the titles and abstracts of all records to identify potentially eligible studies. Then, a full-text review was performed to determine final inclusion according to the eligibility criteria. In cases of disagreement, a consensus was reached by discussion.The following criteria were used to identify eligible studies: (1) selected participants with psoriasis diagnosed using specific criteria and controls without psoriasis; (2) used case–control or cohort study designs; (3) provided sufficient information to calculate OR or hazard ratio (HR) and a 95% CI; and (4) published in English. Exclusion criteria included (1) conference or meeting abstracts, summaries, reviews, comments, letters, news, case reports, case series, and editorials; (2) studies without appropriate data; (3) studies that did not confirm whether drugs were taken before diagnosis; (4) studies not medically classified as psoriasis; or (5) repeated study population.The following information was extracted from each study: name of the first author, publication year, country, study design, definition of patients with psoriasis, the number of patients with psoriasis , the population size , OR or HR with 95% CI, and comorbidities when available.32. The scoring system of this scale was based on three components: subject selection (0–4 points); comparability of study groups (0–2 points); and exposure (case–control study) or outcome . The highest NOS score available for each publication was 9 points.NOS was used to assess the quality of selected studiesp value < 0.05 was considered statistically significant. It was assumed that the OR calculated from case–control studies approximates the HR in cohort studies. Heterogeneity was assessed using the Q test and I2 test33. Based on the heterogeneity results, either the fixed-effects model (Mantel–Haenszel method) or the random-effects model  was used to calculate effect size35.We used ORs with 95% CIs as measures of effect size of ACE inhibitors for psoriasis. Statistical significance was analyzed using the z test, and a 36. A p value < 0.05 was considered statistically significant. Sensitivity analysis using the leave-one-out method was performed to assess the stability of results. Subgroup analyses by study design, continent and climate were also conducted if there were at least two studies in each subgroup. All statistical analyses were performed using Review Manager (RevMan) version 5.4  and R software . The paper was written according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines37.The Egger regression test of the funnel plot was performed to identify publication bias