ALK+ NSCLC.The discovery of specific oncogenic drivers in non–small cell lung cancer (NSCLC) has led to the development of highly targeted anaplastic lymphoma kinase tyrosine kinase inhibitors (ALKis). Brigatinib is a next-generation ALKi associated with prolonged progression-free survival in patients with ALKi-naive ALK+ NSCLC, from a US payer perspective.To estimate the cost-effectiveness of brigatinib compared with crizotinib and alectinib in patients with ALKi-naive ALK+ NSCLC setting. Brigatinib was compared with crizotinib within a cost-effectiveness framework and compared with alectinib in a cost-comparison framework, where all efficacy outcomes were assumed equal. The efficacy of brigatinib and crizotinib was informed by the ALTA-1L trial, and an indirect treatment comparison was performed to inform the efficacy of brigatinib vs alectinib owing to a lack of head-to-head data. Costs were derived from public sources. The main outcomes of the model were total costs, quality-adjusted life-years (QALYs), life-years, and incremental cost-effectiveness ratios. Univariate and probabilistic sensitivity analyses, in addition to multiple scenario analyses, were conducted to assess the robustness of the model outcomes.A lifetime area under the curve–partitioned survival model with 4 health states was used to evaluate the relative cost-effectiveness of brigatinib in the ALKi-naive The improved outcomes observed in ALTA-1L translated into QALY gains (+0.97) in the comparison of brigatinib vs crizotinib. The superior efficacy profile was associated with increased time on treatment with brigatinib, which drove the increase in costs vs crizotinib . The resulting base-case incremental cost-effectiveness ratio was $217,607/QALY gained. Compared with alectinib, brigatinib was associated with a cost difference of −$8,546. Sensitivity analysis suggested that extrapolation of overall survival, the assumptions relating to time on treatment, and subsequent therapy costs were the most influential determinants of results. Probabilistic sensitivity analysis suggested brigatinib had the highest probability of being cost-effective beyond willingness-to-pay thresholds of $236,000 per QALY vs crizotinib and alectinib.At list prices and under base-case assumptions in the current analysis, brigatinib was associated with cost-savings vs alectinib, and QALY gains but at higher costs vs crizotinib. Additional research into the real-world efficacy of ALKis is warranted to further understand the comparative cost-effectiveness of these therapies. This study compares 3 lung cancer treatments in the United States. The results show that treatment A (brigatinib) is associated with a higher quality of life but higher costs than treatment B (crizotinib). Treatment A is associated with the same quality of life but lower costs than treatment C . Further research is needed to understand how these lung cancer treatments affect patient well-being and costs. Implications for managed care pharmacyThis is the first published economic evaluation comparing anaplastic lymphoma kinase inhibitors  for the treatment of ALK + non–small cell lung cancer. Base-case results demonstrated variable quality-adjusted life-years and costs in a US setting. A greater understanding of the cost-effectiveness of US Food and Drug Administration–approved ALKis could better inform clinical trial investment, formulary, and US policy.2 Non–small cell lung cancer (NSCLC) is the most prevalent histologic class of lung cancer, accounting for approximately 85% of cases.4 Anaplastic lymphoma kinase (ALK) is a tyrosine kinase that is aberrantly expressed in 3%-7% of patients with NSCLC (referred to as ALK+ NSCLC).6 The presence of chromosomal rearrangements involving the ALK oncogene defines a molecular subset of NSCLC with distinct demographic features, including earlier age of onset and a higher prevalence in women.8 For patients diagnosed with stage IV ALK+ NSCLC, median overall survival (OS) is 81 months .9Lung cancer is one of the most commonly diagnosed cancers in the United States, accounting for 12.4% of all new cancer cases.10 Crizotinib was the first ALKi to be approved for the treatment of metastatic ALK+ NSCLC by the US Food and Drug Administration (FDA) in 2011.11 Since then, a series of next-generation ALKis has been approved by the FDA for the frontline treatment of metastatic ALK+ NSCLC: ceritinib (2017),12 alectinib (2017),13 brigatinib (2020)14 , and lorlatinib (2021)15 (third-generation). Brigatinib is a central nervous system (CNS)-active, next-generation ALKi that targets a broad range of ALK-acquired resistance mutations. The ALTA-1L head-to-head trial of brigatinib vs crizotinib demonstrated that brigatinib is associated with prolonged progression-free survival (PFS) in patients with ALKi-naive ALK+ NSCLC.17The discovery of specific oncogenic drivers in NSCLC has led to the development of highly targeted ALK tyrosine kinase inhibitors (herein referred to as ALK inhibitors [ALKis]).ALK+ NSCLC are at risk of developing CNS metastases. CNS metastases are present in 22%-33% of patients at initial diagnosis and are the most common relapse site following ALKi treatment (45%-70%).22 CNS involvement can cause physical and psychological impairments and is typically managed with surgery, radiotherapy, and/or systemic therapy.23 CNS metastases remain a major cause of mortality among patients with ALK+ NSCLC, with only 4% of patients surviving for 5 or more years.24 CNS metastases are also associated with burdensome symptoms and a low health-related quality of life (HRQoL).24 Owing to CNS metastases constituting a considerable clinical burden and influencing treatment decisions, the consequences of CNS and non-CNS progression were captured separately in the economic model.Many patients with 25 Analysis of 2 US claims databases demonstrated that substantially higher costs are incurred for patients with ALK+ NSCLC with CNS metastases vs patients without CNS metastases; mean total unadjusted costs per patient per month were $29,497 vs $22,791, respectively .26It is estimated that US expenditure for lung cancer care was more than $14.1 billion in 2018.ALK+ NSCLC,27 the disease remains incurable and annual lung cancer deaths (85% of which are NSCLC) in the United States are still greater than the combined deaths from colorectal, pancreatic, and breast cancers.4 The approval of multiple ALKis in the frontline setting has improved the treatment options available to patients.28 However, each ALKi has its own risk/benefit profile, and patients ultimately progress.30 Therefore, there is a continued need for novel efficacious frontline therapies to improve the patient journey and treatment pathway.Although advances in targeted therapies have improved survival in patients with ALK+ NSCLC, from a US payer perspective. These comparators were considered most appropriate as alectinib was considered the market-leading ALKi in the first-line setting at the time of analysis, and head-to-head data for brigatinib vs crizotinib were available from the ALTA-1L trial.29 Although a cost-effectiveness analysis of brigatinib vs crizotinib was possible because of the availability of head-to-head data,29 a cost comparison was undertaken vs alectinib owing to results from an indirect treatment comparison (ITC) being near unity, and following clinical opinion that brigatinib provides similar health benefits to alectinib. This study is the first economic evaluation comparing brigatinib, crizotinib, and alectinib for the treatment of ALK+ NSCLC in the United States.The objective of this study was to estimate the cost-effectiveness of brigatinib vs crizotinib and a cost comparison of brigatinib vs alectinib in patients with ALKi-naive ALK+ advanced NSCLC setting.31 The model was developed in Microsoft Excel 2010 as an area under the curve–partitioned survival model with 4 health states: preprogression, CNS progression, non-CNS progression, and death .29 Owing to the lack of head-to-head data, an ITC was performed to inform the efficacy of brigatinib vs alectinib. Based on the results of the ITC, in which all results were near unity and nonsignificant, and following clinical feedback that suggested clinicians considered brigatinib to provide similar health benefits to alectinib, brigatinib was compared with alectinib in a cost-comparison framework. Within this framework, PFS, CNS-PFS, and OS for alectinib were assumed equal to those observed in the ALTA-1L trial for brigatinib .The model estimated the cost-effectiveness of brigatinib vs crizotinib and a cost comparison of brigatinib vs alectinib in patients with ALKi-naive 33The model considered a lifetime horizon (30 years) and a 28-day cycle length. Discounting was applied at 3% to the costs, quality-adjusted life-years (QALYs), and life-years gained in the base case.17 This is anticipated to reflect clinical practice in the United States, where brigatinib is available after crizotinib.23 Therefore, no adjustments were made in the base case. Scenario analyses considered the impact of removing efficacy and costs relating to subsequent brigatinib from the crizotinib arm using treatment switching methods.In the ALTA-1L clinical trial, patients who experienced progression as assessed by the blinded independent review committee (BIRC) or received radiotherapy to the brain while on crizotinib therapy could cross over to treatment with brigatinib. A total of 65 patients (47.1%) in the crizotinib arm formally crossed over to brigatinib. However, further informal crossover  was also observed; in total, 80 patients (58.0%) in the crizotinib arm received subsequent brigatinib.29 Systematic literature reviews, supplemented with targeted literature searches, identified the relevant data for the ITC with alectinib. The ITC was informed by 3 publications for alectinib vs crizotinib from the ALEX trial.35The efficacy for brigatinib and crizotinib was informed by the final data-cut from the ALTA-1L trial (median follow-up 40.4 months).Owing to the differences in a key treatment effect modifier (baseline CNS metastases) between the clinical trials, population-adjusted indirect methods  were considered with baseline CNS metastases included as the only treatment effect modifier, in addition to the standard network meta-analysis methodology.36The OS data in the crizotinib arm from the ALTA-1L clinical trial reflect a sequence of treatments (crizotinib followed by brigatinib), due to the high rate of crossover. Therefore, several methods were explored to adjust for the impact of crossover in the ALTA-1L clinical trial within the ITC framework, including inverse probability of censoring weighting, marginal structural models, and rank-preserving structural failure time models . These adjusted data were used in the ITCs to obtain relative efficacy estimates free of confounding from crossover in the crizotinib arm of the ALTA-1L trial. Full details of the methodology are reported elsewhere.35 Clinical experts also communicated that brigatinib provides similar health benefits to alectinib. Thus, a cost-comparison analysis was conducted for brigatinib vs alectinib in the base case, in which PFS, OS, and CNS-PFS outcomes were assumed equivalent. A scenario analysis considered a cost-effectiveness framework for this comparison using the point estimates from the ITC. Additional scenarios explored the ITC methodologies, treatment switching adjustments, and the assessment body for PFS .The ITCs indicated similar outcomes under alternative assumptions for brigatinib vs alectinib for PFS BIRC (range for hazard ratios: 0.946-1.036) and OS (range for hazard ratios: 0.902-1.088). No ITC was possible on CNS-PFS outcomes because of differences in endpoint definitions in the ALTA-1L and ALEX trials.17 Assessment of proportional hazards was used to determine whether to use stratified or independent parametric models; clinical feedback indicated that because of the different mechanisms of action, they would not expect proportional hazards to hold between brigatinib and crizotinib. Therefore, independent parametric models were fitted to brigatinib and crizotinib data. However, for indirect comparisons, clinicians would expect proportional hazards to hold between brigatinib and alectinib based on similar mechanisms of action across the second-generation/third-generation ALKis.To inform the inputs for brigatinib and crizotinib in the model, data from ALTA-1L were extrapolated for the following outcomes: OS, PFS BIRC, and CNS-PFS .17 AssesSeven parametric distributions  were fitted to the patient-level data for each outcome. The fit of each parametric model to the survival data was assessed through Akaike information criterion/Bayesian information criterion statistics, comparison with the Kaplan-Meier curves, and clinical expert opinion.For all outcomes, there were limited differences in terms of how well each parametric curve fitted the observed data. The exponential curve was selected in the base case for both brigatinib and crizotinib for all outcomes. This represented the most conservative curve , and the selection was validated by clinical experts.37 and the lung cancer–specific module .38 European Organization for Research and Treatment of Cancer quality of life questionnaire data were mapped to European Quality of Life-five dimensions-3 levels utility scores (US tariff)39 using the best fitting response mapping algorithm published in Longworth et al.40 A mixed-effects regression model, to account for repeated measures, was then fitted to the data to estimate utility values for the different health states. Prognostic factors in the regression analysis included the presence of grade 3/4 adverse events (AEs), baseline European Quality of Life-five dimensions-3 levels index score, complete response, partial response, or stable disease (vs progressive disease). To reflect HRQoL throughout the model time horizon, and beyond the follow-up from ALTA-1L, multipliers sourced from the literature were applied accounting for the impact of CNS progression, chemotherapy, and best supportive care . In addition, the model included a utility decrement associated with increasing age (−0.00026 each year).41HRQoL was collected in ALTA-1L until 30 days after the last study dose with the European Organization for Research and Treatment of Cancer quality of life questionnaire (version 3.0)Supplementary Table 5).44 In the base case, dose interruptions/reductions associated with AEs or noncompliance were accounted for. The median relative dose intensities were 94.68% for brigatinib, 99.08% for crizotinib, and 100% for alectinib.45 Subsequent therapies were based on US clinical practice. Scenarios are presented that assume subsequent therapy use as per ALTA-1L (Supplementary Tables 2 and 8) and exclude the use of dose intensity calculations.All pharmacy costs were obtained from the Micromedex REDBOOK, and dosing schedules were aligned with marketing authorizations (Supplementary Table 7).Any-cause grade 3/4 AEs occurring in greater than or equal to 3% of patients in ALTA-1L and ALEX were included in the analysis. AEs were only modeled for patients on treatment and it was assumed that AEs for all therapies ceased once treatment was discontinued. It was further assumed that AEs lasted 1 model cycle (28 days). Where AEs were not reported in specific publications, the model assumed an event rate of 0. The costs associated with laboratory abnormalities were assumed to include the cost of 2 medical oncology outpatient visits and 2 blood tests (46 and associated costs were sourced from the Centers for Medicare and Medicaid Services Physician Lookup47 (Supplementary Tables 4 and 6).Health state resource use was defined by whether a patient received frontline treatment , had discontinued frontline treatment , or had CNS progression. Resource use inputs were informed by Zhou et al,48The cost of end-of-life care was applied to all patients who entered the death health state as a one-off cost. The one-off cost of terminal care was $1,863, calculated as the average management cost associated with the 8 weeks prior to death, including hospitalization, hospice care, and outpatient patient costs.One-way sensitivity analyses were performed to evaluate the sensitivity of model results to individual inputs. Each parameter was independently varied within a plausible range defined by the 95% CI or a credible interval. The results for the most influential parameters were plotted on a tornado diagram. Scenarios exploring the following assumptions were also performed: population, parametric curve fits for OS, PFS BIRC, PFS investigator-assessed, CNS-PFS, and treatment switching adjustments. To characterize joint uncertainty in model inputs, a probabilistic sensitivity analysis (PSA) was performed. A total of 10,000 PSA iterations were performed and results plotted on a cost-effectiveness acceptability curve, which reports the probability that each therapy is the most cost-effective treatment option at alternative willingness-to-pay (WTP) thresholds.Supplementary Figure 1) demonstrated that the cost of subsequent therapies was also a key driver of results. A scenario exploring the use of subsequent therapies based on those observed in ALTA-1L reduced the ICER to $165,990/QALY gained. Further scenarios exploring subsequent therapy with ceritinib at a dose of 750 mg and assuming 100% dose intensity for crizotinib led to ICERs of $215,138/QALY gained and $240,176/QALY gained, respectively. Key base-case assumptions relating to the comparison of brigatinib vs crizotinib are conservative, as alternative assumptions generally improved the ICER in favor of brigatinib (Supplementary Table 2).As reflected in the scenario analyses, the ICER was most sensitive to the choice of OS parametric curve and the assumptions relating to time on treatment. The exponential curve predicted OS outcomes in the base case; this is a conservative assumption because it was associated with the highest ICER, as other parametric forms reduced the ICER as low as $169,483/QALY gained (Gompertz). The base case assumed patients receive treatment until progression—again, this is considered a conservative assumption because using the time on treatment data from the ALTA-1L clinical trial reduced the ICER to $98,436/QALY gained. Furthermore, alternative assumptions relating to duration of treatment led to ICERs that ranged from $211,011/QALY gained (treat for 3 cycles after progression) to $217,607/QALY gained (treatment until progression). ICERs for scenarios exploring treatment switching adjustments ranged from $215,143/QALY gained  to $246,051/QALY gained . One-way sensitivity analysis (Supplementary Table 2).Compared with alectinib, brigatinib was associated with a cost difference of −$8,546 attributable to the subsequent therapy distributions . In the scenario that sourced subsequent therapy distributions from ALTA-1L and relevant clinical trials, brigatinib was associated with a cost increase of +$72,740. Treatment duration scenarios led to incremental cost savings that ranged from −$8,338 (treat for 3 cycles after progression) to −$8,477 (treat for 1 cycle after progression). Assuming 100% dose intensity for alectinib led to a marginal cost increase with brigatinib of +$21,743 (Supplementary Table 3). However, these should be interpreted carefully as the very small difference in incremental QALYs—driven by the limited differences in efficacy between the 2 drugs—results in an inflated ICER.In the cost-effectiveness scenarios, ICERs for brigatinib vs alectinib ranged from $21,524/QALY gained to $304,461/QALY gained using different ITC and treatment switching methodologies. This included ICERs of $92,898/QALY gained under an anchored MAIC  adjusting for treatment switching in all switchers, without recensoring scenario; $122,218/QALY gained under an anchored MAIC  adjusting for treatment switching in all switchers and including recensoring scenario; and $149,298/QALY gained under an unanchored MAIC  scenario .Within PSA based on list prices, brigatinib was associated with the highest probability of being cost-effective at all WTP thresholds above $236,000 , which are partly offset by subsequent therapy and terminal care cost savings with brigatinib.The ALTA-1L head-to-head trial demonstrated prolonged PFS with brigatinib vs crizotinib, translating to additional discounted life-years of 1.01 and additional QALYs of 0.97 within the model.ALK+ NSCLC setting, is associated with cost savings of −$8,546 per patient attributable to the subsequent therapy distributions. Although time on frontline treatment and the distribution of patients progressing on brigatinib or alectinib are expected to be similar, patients would not be expected to receive the same treatment again. Therefore, the subsequent therapy distribution in the model was assumed to be equal, with the exception of 30% of patients receiving alectinib after brigatinib and 30% of patients receiving brigatinib after alectinib. Although highly sensitive to subsequent therapy assumptions, alectinib is associated with higher subsequent therapy costs, as the pack price of brigatinib is greater than that of alectinib . Additionally, time on treatment with second-line brigatinib is longer than that with second-line alectinib, as brigatinib has been shown to be highly efficacious in this setting .51 Assuming a dose intensity of 100% for brigatinib led to brigatinib being associated with a marginal cost increase of +$21,743 vs alectinib, further illustrating that cost savings with brigatinib are dependent on drug acquisition costs.The cost comparison of brigatinib vs alectinib demonstrated that brigatinib, in the ALKi-naive 47 The ease of brigatinib dosing offers greater convenience compared with other ALKis owing to its flexibility and reduced pill burden, which could positively impact adherence and convenience for patients, many of whom are of working age.There are some important considerations regarding treatment with brigatinib that are not reflected within the model. Brigatinib is administered as a single tablet once daily with or without food. Therefore, brigatinib provides dosing advantages compared with alectinib (administered as 4 capsules twice daily with food) and crizotinib (administered twice daily).22 Indeed, crizotinib has limited efficacy in the treatment of CNS metastases.52 However, next-generation ALKis  have greater diffusion across the blood-brain barrier, thereby reducing CNS involvement.54 Thus, there are considerable clinical and economic benefits of treating patients upfront with the best possible ALKi.49Many patients treated with crizotinib progress within 2 years, with the CNS being the most common relapse site.29 In addition to efficacy and safety results, ALTA-1L also provided HRQoL data for analysis. It is noteworthy that clinician feedback suggests that this trial is highly representative of real-world clinical practice, including prior systemic chemotherapy, baseline disease characteristics , and allowing crossover from crizotinib to brigatinib. The data from ALTA-1L and ALEX were based on final and latest datacuts, respectively, for each trial. Differences between study populations used in the ITC were explored using a MAIC methodology. Additional strengths of the analysis include the model structure reflecting outcomes of importance to patients and clinicians  and extensive sensitivity analyses being conducted to explore the assumptions and uncertainty associated with different data sources and methods.A key strength of this analysis is the robust direct comparison between brigatinib and crizotinib in the phase 3 ALTA-1L trial.55—other parametric curve selections were found to overpredict long-term survival.Following extensive validation of outcomes by clinical experts, the exponential curves for OS, PFS, and CNS-PFS were the only plausible extrapolations that provided internal consistency for these outcomes within the model. The exponential distribution was the most conservative curve choice for OS outcomes for brigatinib vs crizotinib. The predicted life-years from the exponential curves also align with published economic evaluations and literature24The analysis is subject to several limitations. CNS-PFS data are not currently available for alectinib; given that CNS metastases are a major cause of mortality and are associated with a low HRQoL, the analysis would benefit from the inclusion of these data.17 whereas the ALEX trial defined a PFS event as a Response Evaluation Criteria in Solid Tumors progression or death, whichever occurred first.35 The number of events defined by radiotherapy to the brain was small in the ALTA-1L trial: 3 (2.2%) and 9 (6.5%) in the brigatinib and crizotinib arms, respectively, for PFS BIRC.29 Thus, although the impact of this on PFS BIRC, PFS, and CNS-PFS outcomes should be considered, it was not thought to be a driver of results. In addition, the base case assumed that patients are treated until progression; in a real-world clinical setting, patients may be treated beyond progression, which reduces the ICER in favor of brigatinib.The ALTA-1L trial defined a progression event as a Response Evaluation Criteria in Solid Tumors progression, radiotherapy for CNS metastases, or death, whichever occurred first,There are incomplete data from both trials regarding the subsequent therapies received after frontline therapy. The unadjusted OS data reflect a sequence of treatments, with the effects of the subsequent therapies confounding relative efficacy estimates of frontline therapies. The crizotinib arm in the ALTA-1L study was particularly impacted by this owing to the subsequent use of brigatinib in addition to other subsequent ALKi use. Although methods were explored to disentangle the effects of crossover using the ALTA-1L data, it was not possible to do this for other subsequent therapies given the small patient numbers supporting the complex methodology and the lack of information from the ALEX trial. The authors are unable to comment on which direction this would drive results.17 Brigatinib incurred higher costs than crizotinib, mostly explained by improved efficacy with brigatinib, which resulted in patients remaining progression-free and receiving treatment for longer. The cost comparison of brigatinib vs alectinib assumed OS, PFS, and CNS-PFS outcomes were equal and demonstrated that brigatinib was cost-saving under base-case assumptions. Additional research into the real-world efficacy of ALKi therapy is warranted to further understand comparative cost-effectiveness.The ALTA-1L head-to-head trial demonstrated prolonged PFS with brigatinib vs crizotinib, translating to additional life-years and QALYs within the economic model.


Social motivation—the psychobiological predisposition for social orienting, seeking social contact, and maintaining social interaction—manifests in early infancy and is hypothesized to be foundational for social communication development in typical and atypical populations. However, the lack of infant social-motivation measures has hindered delineation of associations between infant social motivation, other early-arising social abilities such as joint attention, and language outcomes. To investigate how infant social motivation contributes to joint attention and language, this study utilizes a mixed longitudinal sample of 741 infants at high (HL = 515) and low (LL = 226) likelihood for ASD. Using moderated nonlinear factor analysis (MNLFA), we incorporated items from parent-report measures to establish a novel latent factor model of infant social motivation that exhibits measurement invariance by age, sex, and familial ASD likelihood. We then examined developmental associations between 6- and 12-month social motivation, joint attention at 12–15 months, and language at 24 months of age. On average, greater social-motivation growth from 6–12 months was associated with greater initiating joint attention (IJA) and trend-level increases in sophistication of responding to joint attention (RJA). IJA and RJA were both positively associated with 24-month language abilities. There were no additional associations between social motivation and future language in our path model. These findings substantiate a novel, theoretically driven approach to modeling social motivation and suggest a developmental cascade through which social motivation impacts other foundational skills. These findings have implications for the timing and nature of intervention targets to support social communication development in infancy. This study takes a dimensional approach in a large, heterogeneous sample of infants at high and low familial likelihood of ASD to examine associations between continuously distributed domains of functioning across the spectrum of typical to atypical development .We hypothesized that items spanning all the theorized social-motivation elements  from common parent-report questionnaires would together represent a single latent factor across 6 and 12 months of age. To characterize more accurate developmental relations, we tested and adjusted for measurement non-invariance, which we explored for age, sex, and familial likelihood. We hypothesized cascading effects evidenced by positive longitudinal associations between 6-month social motivation, as well as growth in social motivation from 6 to 12 months, and IJA and RJA, both measured at 12–15 months of age. Additionally, we hypothesized positive relations between 6-month social motivation and growth in social motivation with receptive and expressive language outcomes at 24 months of age. Based on prior work, we also hypothesized positive relations between RJA, IJA, and 24-month language.4 |4.1 |Participants came from the Infant Brain Imaging Study (IBIS), a prospective longitudinal study of infant siblings at high familial risk, and therefore high likelihood (HL), for ASD by virtue of an older sibling with ASD, as well as low likelihood (LL) infant siblings with an older sibling without ASD . The incData collection occurred at four study sites: University of North Carolina at Chapel Hill; Children’s Hospital of Philadelphia; Washington University in St. Louis; and University of Washington. Analyses included a mixed longitudinal sample of 741 infants (515 HL infants and 226 LL infants). Four hundred and twenty infants contributed three time points of data , 218 infants contributed two time points of data, and 103 infants contributed only one time point. Trained clinicians provided a clinical best estimate diagnosis at 24 months for all infants based on assessment which included the Autism Diagnostic Observation Schedule ADOS;  and revi4.2 |4.2.1 |The IBIS battery was reviewed for items querying social behavior in established parent-report instruments (those considered are described below) which index individual differences in infancy and represent parental impressions of behavior across time and contexts. Candidate social-motivation items, collected at 6 and 12 months, were selected based on face validity for As described in a separate manuscript from our group validating an initial series of items in a social motivation composite , item selection was initiated by the senior author, who selected a preliminary series of parent-report items as candidate indicators of social motivation from available 6- and 12-month measures. At 6 months, parent-report items were considered from the Vineland Adaptive Behavior Scales VABS; , a measuFor the 12-month time point, in addition to the VABS and IBQ-R, parent-report items were considered from the Macarthur-Bates Communicative Development Inventory MB-CDI; , a measuCandidate items at this stage were next tested for internal reliability of these preliminary items using Cronbach’s alpha, and items lowering internal reliability at 6 or 12 months were eliminated. A second face validity check was performed by three co-authors  who reached a consensus, resulting in 22 6-month indicators and 40 12-month indicators. Items were again reviewed for face validity by the first author of the present study. As a preparatory step for MNLFA, a series of single-factor CFA models were conducted across groups divided by sex, likelihood, and age. Standardized loadings accommodated the different scales of the items. The final 23 items for this study were identified based on good model fit in the context of high loadings on a single social motivation factor, maximized overlap across 6- and 12-month items, and sufficient indicator-level variation (see 4.2.2 |Initiating joint attention (IJA) was measured by coding the infant’s acts used to direct another’s attention based on the Communication and Symbolic Behavior Scales–Developmental Profile behavioral assessment CSBS-DP; , which iResponding to joint attention (RJA) was measured using the Dimensional Joint Attention Assessment DJAA; , a play-4.2.3 |Twenty-four-month language outcomes were measured using the receptive and expressive subscale t-scores of the Mullen Scales of Early Learning MSEL; . The MSE4.3 |4.3.1 |lavaan package  models in R Studio using the  package  to identnon-invariant, or exhibit measurement bias, if it functions differently for different groups of participants at a given level of the latent variable.Next, to establish MI for the indicators of latent social motivation derived from the previous step, we used moderated nonlinear factor analysis MNLFA; . Items iaMNLFA  by a specified set of covariates  covariates moderate factor means and variances as well as item intercepts and factor loadings in the latent factor model  and DIF (b and c) effects simultaneously to form the final scoring model using the significant effects established in the previous steps and implements Benjamini-Hochberg correction for multiple comparisons to create the most parsimonious model. Finally, it obtains parameter estimates for all significant sources of DIF to create a final latent social-motivation scoring model that then generates social-motivation factor scores for each infant in the entire sample at 6 and 12 months of age. We assessed mean impact by familial likelihood, age, and sex; variance impact by age; and included HL, age, and sex as covariates  framework, as it allowed us to simultaneously model multiple dependent variables, estimate latent variables, model residual covariance, and capitalize on full information maximum likelihood (FIML) to accommodate missing data. For instance, with this framework, we can account for known relations between joint attention and language in addition to the relations with social motivation. Standardized beta weights are based on the variances of both the observed and latent variables.4.3.3 |Missing data were assumed to be missing at random (MAR)—that is, conditionally missing, given the other variables included in the observed covariance matrix . A stren5 |5.1 |5.1.1 |CFAs fitted to the different groups of age, sex, and familial likelihood status found that 23 items loaded sufficiently onto a single latent social-motivation factor for all groups . Fit statistics by group are shown in 5.1.2 |aMNLFA revealed significant DIF, or measurement non-invariance, for 11 of the social-motivation indicators spanning the different social-motivation aspects, as shown in B = −0.274, SE = 0.096, p = 0.004) and age . This suggests that, on average, HL infants exhibited lower = − social-motivation scores compared to LL infants and social-motivation scores increased with age, after adjusting for all DIF. The final aMNLFA scoring model can be found in 5.2 |r = 0.61. The final SEM model simultaneously modeling relations between social motivation, RJA , IJA , receptive language , and expressive language  is shown in p < 0.001; Bstd = 0.63; as indicated by model intercept values). A negative relationship between 6-month social motivation and 6–12 month growth in social motivation was observed, in which, on average, higher 6-month social motivation was associated with lower 6–12 month growth, and lower 6-month social motivation was associated with higher 6–12 month growth .Six and 12-month social-motivation values were correlated at p = 0.004; Bstd = 0.12) and RJA by 0.19 units  measured between 12–15 months, regardless of social-motivation levels at 6 months of age. We also found significant positive associations between IJA and both receptive  and expressive  language, as well as between RJA and both receptive  and expressive  language.A significant relationship between 6- to 12-month growth in social motivation and IJA was observed. On average, infants who exhibited greater increases in social motivation from 6 to 12 months demonstrated significantly greater IJA by 0.15 units (p’s > 0.55). Additionally, there were no statistically significant relations between 6 or 12-month social motivation and either joint attention or language (p’s > 0.45).This model showed no statistically significant associations between social-motivation growth and either receptive or expressive language at 24 months of age  and receptive  language abilities when joint attention was omitted from the original model.A secondary analysis, in which sex was added as a covariate, showed that all identified relationships remained significant, without substantive changes in magnitude . An expl6 |Early social motivation has been hypothesized to provide a foundation for social communicative development, although this notion has been difficult to test given the challenge of establishing measures of social motivation in infancy. The advancements from the present study are twofold: (1) we demonstrate the feasibility of a latent model of social motivation in the first year of life, drawing on items from common parent-report measures, that exhibits measurement invariance for a heterogenous sample of infants; and (2) we provide some evidence for cascade effects for the development of social motivation in infancy and future joint attention abilities for HL and LL infants.6.1 |This study extends past work identifying indicators of social motivation in infancy . We founSimilar to other longitudinal work examining constructs early in development e.g., , we foun6.2 |In partial support of our hypotheses, we found some evidence for cascade effects, or those spreading over time and domain, for early social motivation. Infants who exhibited greater intra-individual growth in social motivation between 6 and 12 months of age also exhibited significantly more IJA behaviors at the end of the first year of life, which in turn, were associated with greater 24-month language skills. Growth in social motivation was negatively associated with 6-month social motivation, illustrating that infants with low social motivation exhibited on average greater growth in social motivation than infants with high 6-month social motivation, consistent with an observed upper limit to the distribution of social motivation scores  or potenr = 0.61, a level comparable to established traits including IQ, with r = 0.63  in infants, with r = 0.63  and persr = 0.55 . At thisOf note, the potentially cascading effect of social-motivation growth takes place during a known sensitive period for social brain and behavior development . The secThe importance of growth in social motivation, and its potential role in a developmental cascade in this period, is also consistent with the pre-diagnostic emergence of signs of ASD. Our findings suggest that while growth in social motivation may enhance developmental outcomes, decreases or plateauing of the level of social motivation during this period may contribute to a course of increasingly atypical development. Relative to ASD, While growth in 6- to 12-month social motivation was associated with IJA, we did not confirm our hypothesis for RJA, which showed only a trend-level relation with social motivation. The finding for IJA supports existing theory e.g.,  that in We also did not identify separate, direct associations of social motivation with receptive or expressive language in this developmental model. In an exploratory analysis, we could detect significant associations between 6-to-12-month growth in social motivation and 24-month language measures when joint attention was not included, suggesting that relations between social motivation and language were not significant over and above the relations between joint attention and language, consistent with the role of social motivation as part of a developmental cascade. Lastly, we observed no significant relations between social-motivation levels at 6 months and either joint attention or language abilities. This further suggests that in the case of 24-month language skills, growth in social motivation during the first year of life, with associated cascade effects on IJA, contributes to future language development. This finding underscores the importance of intensive longitudinal designs, rather than only cross-sectional designs, to monitor the course of early development. It also raises the possibility that early surveillance of social motivation could inform opportunities for interventions to stimulate socially rewarding experiences within the first year, which in turn could support the development of joint attention and improve functional communication outcomes. Future work will investigate the systems underlying social-motivation growth in the first year of life, including its neurophysiological, experiential, and genetic underpinnings.7 |Findings from this study should be interpreted in light of several limitations. Given that our social-motivation items were derived from a dataset with different items at each age, some indicators of social motivation differed by age or were unique to a specific age. While using all available indicators maximized detectable variation in social motivation during development, we could not fully disaggregate age-related versus indicator-related changes in social motivation and thus only established partial MI. In our mixed longitudinal design, some infants contributed data at a subset of time points or provided concurrent rather than serial data on social motivation and JA metrics at 12 months. These issues limited our interpretation of directionality of relations, although by maximizing subject inclusion, our approach increased statistical power and precision of model estimates within a more representative sample. Relatedly, our correlational study design does not allow causal claims about the relations between the constructs. Replication in more socio-demographically diverse study populations with dense longitudinal designs and developmentally continuous indicators of social motivation is warranted to clarify relations between social domains and infant developmental outcomes.This study also relied only on parent report, a subjective measurement approach. While parent report is established in child clinical practice (e.g., for language; Lastly, item indicators in our measure could not differentiate the extent to which a behavior with face validity for social motivation reflects an infant’s disposition to engage socially or social skill, and real-world manifestations of social motivation could include both elements. Our leveraging a latent modeling approach allowed us to extract common variation related to social motivation across diverse indicators drawn from developmental questionnaires designed for different purposes. Future psychometric work on social motivation will benefit from incorporating diverse measurement modalities and approaches, including clinician-ascertained and biological indices. Incorporation of a greater array of social metrics than available in our existing dataset will also allow additional validity testing to investigate relative convergence and divergence with other aspects of social behavior, as well as measurement invariance for domains that support general function, such as cognition. Studies in larger atypically developing samples, including those with cognitive delays, will also allow follow-up evaluation of the generalizability of the measurement of social motivation and its developmental associations.8 |In conclusion, this study offers a novel theoretically consistent approach to quantifying social motivation and explicating its role in early development. Findings provide support for a latent model of social motivation based on parent-reports of infants’ tendencies to orient socially and to seek, like, and maintain social engagement within a heterogeneous sample of infants. Developmental modeling adjusting for measurement non-invariance of several indicators by age, sex, and ASD likelihood reveals associations between social-motivation growth during the first year and future joint attention abilities for HL and LL infants. Together, these findings further our understanding of social motivation early in ontogeny with implications for future research investigating the neural bases of early social motivation, its role in ASD, and potential early screening and intervention targets.supplement


Despite a lack of evidence, more than half of patients on dialysis were prescribed statins after an ASCVD event. In patients on dialysis after ASCVD, statin therapy significantly reduced the risk of long-term all-cause mortality.Statin therapy is essential for secondary prevention in patients with atherosclerotic cardiovascular disease (ASCVD). However, the effects of statin therapy in patients receiving chronic dialysis remain uncertain. We aimed to evaluate the effect of statin therapy on long-term mortality in patients on dialysis after a first-time ASCVD. Patients receiving maintenance dialysis aged ≥ 18 years with a first-time ASCVD event between 2013 and 2018 were included in the Korean National Health Insurance Service database. Associations of statin use with long-term mortality were examined using Cox proportional hazards regression models adjusted for demographics and comorbidities. Among 17,242 patients on dialysis, 9611 (55.7%) were prescribed statins after a first-time ASCVD event. Among statin users, 7376 (76.7%) used moderate-intensity statins. During a mean follow-up of 32.6 ± 20.9 months, statin use was associated with a lower risk of all-cause mortality than statin nonuse after adjusting for confounding factors (hazard ratio [HR]: 0.92; 95% confidence interval [CI] 0.88–0.97;  Furthermore, cardiovascular (CV) disease is the leading cause of death in patients receiving dialysis, accounting for more than 50% of deaths in these patients3. Statin, a lipid-lowering agent, is widely used for primary and secondary prevention of ASCVD and is effective in improving CV outcomes and reducing mortality in the general population at risk of ASCVD, including patients with nondialysis-dependent chronic kidney disease (CKD)5. However, recent guidelines8 have not provided any recommendation for the initiation of statin therapy in high-risk patients on dialysis, such as those with preexisting ASCVD, because previous landmark trials have failed to prove the CV benefits11.Patients on dialysis have more atherosclerotic cardiovascular disease (ASCVD) risk factors7. Recently, some observational studies have consistently reported the mortality benefit of statin therapy in high-risk patients on dialysis, such as those with coronary heart disease or a history of coronary revascularization16. A recent retrospective study from the Japan percutaneous coronary intervention (PCI) registry demonstrated that statin therapy was associated with a reduced risk of all-cause death and cardiovascular death, irrespective of the magnitude of the low-density lipoprotein (LDL)-lowering effect12. Two observational studies from the Taiwan National Health Insurance Research Database showed a beneficial effect of statins in reducing the risk of all-cause mortality following myocardial infarction16. In a study of 150 Korean patients requiring chronic hemodialysis who underwent PCI, statin therapy significantly decreased the risk of the composite of nonfatal myocardial infarction, stroke, and all-cause mortality14.Despite the lack of guideline recommendations and insufficient supporting evidence of statin therapy for secondary prevention, statins have been widely prescribed in real-world practice in patients on dialysis who had high-risk profiles, such as those with diabetes or preexisting ASCVD. Additionally, current American and European guidelines recommend continuing statins in patients on dialysis who are already on statins, particularly in those with ASCVDTherefore, we hypothesized that the initiation of statin therapy after ASCVD events might significantly decrease mortality risk in patients on dialysis. This study aimed to examine our hypothesis in patients on dialysis who experienced a first-time ASCVD event using the Korean NHIS  database.17. The NHIS database includes all beneficiaries' medical records, including demographic data, outpatients’ visits, hospitalization, medical history, general health check-ups, procedures, prescribed medications, diagnoses, and mortality. The diagnostic code was based on the International Classification of Diseases, 10th revision (ICD-10). Among adult  of Daejeon St. Mary’s Hospital (IRB number: DC21ZESE0018). The IRB of Daejeon St. Mary’s Hospital waived the requirement for informed consent because the data analyses were performed retrospectively using anonymized data derived from the Korean NHIS database, which is publicly available via the official website (https://nhiss.nhis.or.kr/).This study used the Korean NHIS database. The NHIS is an obligatory health insurance system that covers over 98% of the Korean population, including approximately 50 million people19. The diagnosis of ASCVD was made if the ICD-10 code for each ASCVD component was recorded with a history of admission/emergency department visit at least once or an outpatient clinic visit more than twice. Patients with angina, ischemic stroke, TIA, or PAD were required to have both ICD-10 codes for each diagnosis and additional claim codes such as imaging studies, endovascular procedures, or surgery only to include patients with clinical evidence of ASCVD. We excluded patients with ischemic stroke or TIA if they had a diagnosis of atrial fibrillation on the index date to exclude patients with a cardioembolic cause of stroke. Coronary revascularization was identified using related procedure codes. Other exclusion criteria were patients who (1) died or discontinued health insurance within 30 days of discharge, (2) had a record of a hospital stay longer than one month at index hospitalization, and (3) received renal transplantation before the index date. Finally, a total of 17,242 adult patients with ESRD who had their first ASCVD event were identified as an eligible population for the present study and were then classified into two groups according to whether they received (statin group) or did not receive (no statin group) statin therapy after the ASCVD event. Statin use was defined as a filled prescription during the first 30 days of the index date. Patients with statin use were further divided into four groups based on statin intensity  was used to estimate the statin adherence of the study population. We calculated the PDC by dividing the number of statin prescription days one year after the index date by 365. Patient baseline characteristics were reported on the index date. The Charlson Comorbidity Index (CCI) was calculated to assess patients' underlying medical conditions20. Patients were followed until December 31, 2019, or the date of an exemption from health insurance services or death, whichever occurred first. Detailed information and definitions used in this study are provided in Supplementary Table Patients with ESRD undergoing maintenance dialysis were defined as those who met the following criteria: 1) had ICD-10 codes related to ESRD (N18-19) and (2) had procedure codes of hemodialysis or material codes of peritoneal dialysis fluid at least 3 months from the index date. ASCVD was defined as a composite of coronary heart disease (CHD) , cerebrovascular accident (CVA) ), or peripheral artery disease (PAD) according to the American College of Cardiology and American Heart Association (ACC/AHA) 2013 guidelines had ICD-The study outcomes were all-cause mortality and cardiac mortality during the follow-up period. Mortality data and cause of death were extracted from the death certificate database from Statistics Korea and matched to the medical claim data by Health Insurance and Review Assessment.chi-squared or Fisher’s exact test as appropriate. Differences between groups or continuous variables were analyzed by 2-tailed Student’s t test or the Mann–Whitney U test. The Jonckheere-Terpstra test for continuous dependent variables and Cochran Armitage test for dichotomous dependent variables were used for trend analysis. A multivariable Cox proportional hazards regression model was used to assess the association between mortality and statin use, adjusted for age, sex, index year, and CCI. The Cox proportional assumption was validated with a log–log plot and goodness of fit test, which satisfied this assumption. The effect of statin therapy [reference: no statin group] was expressed as hazard ratios and 95% confidence intervals. In a sensitivity analysis, we evaluated the association between statin use and mortality according to ASCVD type or statin intensity. Subgroup analyses were performed according to age group (ages < 65 years and ≥ 75 years), sex, presence of comorbidities, and CCI. The overall survival rates were compared using a Kaplan–Meier curve, and the log-rank test was used to compare the groups. A 2-sided p < 0.05 was considered to be statistically significant. Statistical analyses were performed using SAS software V.9.4 .Continuous variables were expressed as means Table entclass1pt{minimap = 0.032), whereas statin use showed a marginally lower incidence rate of all-cause mortality in patients with CVA (p = 0.0059) . In patients with CVA, the statin group showed a 12% risk reduction of all-cause mortality compared with the no statin group . In patients with CHD, statin therapy showed a similar trend in reducing all-cause mortality but was not statistically significant . In contrast to all-cause mortality, the statin group showed a higher cardiac mortality rate than the no-statin group in both univariable and multivariable Cox regression models. However, no significant association was observed between statin therapy and cardiac mortality in dialysis patients with each ASCVD subtype.Table p = 0.025; moderate-intensity, fully adjusted HR: 0.91, 95% CI 0.87–0.96, p = 0.0002). In patients with CVA, the moderate-intensity statin group showed a 13%  reduction in all-cause mortality compared with the no-statin group. However, in patients with PAD, the high-intensity statin group showed a significantly higher all-cause mortality . The high-intensity statin group showed a statistically significant higher cardiac mortality than the no statin group in the overall  and CHD cohorts . Low- and moderate-intensity statin or statin/ezetimibe combination therapy showed no significant association with cardiac mortality in the univariable and multivariable Cox regression models.We further examined the association between statin therapy and mortality stratified by statin intensity Table . Low- orSubgroup analysis for all-cause mortality is presented in Fig. In the present study, we investigated the association of statin therapy with long-term mortality in patients on dialysis after a first-time ASCVD event. The main findings of our study can be summarized as follows. First, over half of the patients on dialysis (55.7%) were prescribed statins after a first ASCVD event, with most patients receiving moderate- or high-intensity statins (76.7% of patients taking moderate-intensity statins and 15.6% of patients taking high-intensity statins). Second, statin therapy independently reduced the risk of all-cause mortality in patients on dialysis after a first-time ASCVD event for 32.6 11. Based on previous landmark RCTs11, the Kidney Disease Improving Global Outcomes (KDIGO) 2013 lipid management guideline suggested neither initiating nor discontinuing statins in patients on dialysis for the primary prevention of CV disease8. More recently, the 2018 American College of Cardiology/American Heart Association (ACC/AHA) and 2019 European Society of Cardiology (ESC) and European Atherosclerosis Society (EAS) lipid management guidelines provided no recommendation regarding statin therapy for patients on dialysis, particularly those with ASCVD7. However, the previous three landmark RCTs could not determine the effectiveness of statin therapy in patients on dialysis after ASCVD because these RCTs included only a small percentage of patients on dialysis who had a history of ASCVD. The negative results in previous landmark RCTs were likely influenced by the inclusion of many patients who were unlikely to benefit from statin therapy. Furthermore, post hoc analyses of RCTs demonstrated that statins significantly reduced the risk of cardiac events and death in hemodialysis patients with a high risk of ASCVD, such as those with diabetes21, elevated LDL cholesterol22, or a high burden of atherosclerosis-related CV risk23. Currently, numerous previous studies using real-world data have also demonstrated the beneficial effects of statin use on long-term mortality in patients on dialysis who had high CV risk profiles, such as patients with a history of PCI14 or MI16. These findings implied that the effect of statins in patients on dialysis is heterogeneous, and a specific population with a high burden of atherosclerosis-related CV risk factors, such as patients with diabetes or a history of previous ASCVD, may benefit from statin therapy. Our study also demonstrated that statin use after a first-time ASCVD event in patients on dialysis is beneficial in reducing long-term all-cause mortality, and this beneficial effect of statins is more prominent in patients with diabetes. In contemporary clinical practice, statin prescriptions in patients on dialysis are prevalent despite the lack of guideline recommendations. Among US veterans receiving maintenance dialysis, more than half (58.1%) of patients used statins, and over one-third of statin use was high-dose24, a finding that is consistent with our results. This widespread use of statins in real-world practice, particularly moderate- or high-intensity statins, reflects physicians’ strong urges to prevent future ASCVD events and reduce morbidities and mortality in patients on dialysis.Although statin therapy has proven to have substantial beneficial impacts on reducing ASCVD events and improving mortality in the general population, including patients with CKD, no guidelines exist regarding statin use in patients requiring dialysis with a higher incidence of ASCVD events and CV-related mortality than the general population because previous landmark trials have failed to demonstrate the CV benefit of statin therapy15. Another retrospective cohort study using the Korean insurance data reported that moderate- to high-intensity statins significantly lowered all-cause mortality by 24% but did not influence CV outcomes in patients on dialysis after acute MI16. The following hypotheses could summarize the plausible explanations for the discrepancy between all-cause and cardiac mortality on statin therapy in patients on dialysis. First, the pathogenesis of CV disease in patients on dialysis is complex and markedly different from that in the general population. Additionally, traditional risk factors for ASCVD do not have the same prognostic value in patients with CKD, particularly in patients with advanced CKD, as in the general population. Several nontraditional risk factors, such as volume overload, vascular calcification, oxidative stress, uremic toxins, inflammation, endothelial dysfunction, abnormal lipid modifications, protein-energy wasting, and abnormalities in chronic kidney disease–mineral bone disorder (CKD–MBD), are more prevalent in advanced CKD  and seem to play a far more critical role in the pathogenesis of ASCVD and CV mortality than in the general population25. The influences of these nontraditional factors should account for competing risks and might outweigh the protective effect of statins in preventing CV disease and decreasing CV mortality in patients on dialysis. Second, the beneficial cardiovascular pleiotropic effects of statins, such as increased nitric oxide bioavailability, antioxidant properties, and inhibition of inflammatory responses, work independently of LDL-cholesterol reduction and may reduce non-CV mortality in patients on dialysis26. Therefore, the beneficial effect of statins may not be primarily associated with reduced CV mortality in patients on dialysis.One noteworthy finding of our study was that statin use was independently associated with a reduction in all-cause mortality but not cardiac mortality in patients on dialysis after a first-time ASCVD event. Previous observational studies conducted on patients requiring dialysis with a very high CV risk also showed results consistent with our study. A Taiwan nationwide cohort study demonstrated that statin therapy showed neutral effects on composite CV outcomes but significantly reduced the risk of all-cause mortality in patients with type 2 diabetes mellitus on dialysis after acute MI27. Unfortunately, NHIS data do not capture information about the time since the onset of dialysis for the individual patient. Fourth, the prescribed statin dose or intensity might change over time during the study period. The alternation of statin dose/intensity might influence the effect of statin on long-term mortality. Fifth, statin use before the index date was not considered in the analysis. However, we exclusively included patients who experienced a first-time ASCVD event (a clear indication for statin therapy). Therefore, the effect of statin use before the index date is not expected to be significant. Sixth, we did not evaluate cause-specific deaths, except for cardiac mortality. The potential differential misclassification effect on cardiac mortality might exist in this study. Finally, our results were based on the Korean nationwide health care database, and the limitation of diverse ethnicities can exist for our study. Despite these limitations, a strength of our study includes the large study population on a nationwide scale covering all patients on dialysis in Korea.This study has limitations. First, because of the characteristics of the observational study, potential residual biases from concomitant use of other CV protective medications or unmeasured CV risk factors, such as nutritional status, socio-economic status, accessibility to healthcare service, and smoking status, cannot be ruled out and may have influenced the results. Second, because the data used in this study are based on NHIS data that do not provide laboratory data such as lipid profiles, insufficient lipid-lowering effects with statins may affect the study results. Third, dialysis duration should be considered one of the important competing risk factors. In a study of 14,298 US veterans transitioning to dialysis, patients who recently started dialysis would most likely benefit from statins. However, the benefit diminished as the duration of dialysis became longerDespite a lack of supporting evidence, statin prescription in patients on dialysis after a first-time ASCVD event is prevalent, accounting for over half of patients, and most patients received moderate- or high-intensity statins. Statin therapy might be effective in reducing long-term all-cause mortality in patients on dialysis after ASCVD. However, large scale, well-designed trials are needed to confirm our findings before they can be implemented in clinical practice.Supplementary Tables.


Frailty predicts an increased risk of postoperative morbidity and mortality. Comparison of the predictive performance between two deficit accumulation models of frailty, the modified frailty index (mFI) and the revised-Risk Analysis Index (RAI-rev), is poorly understood. This study compared the predictive abilities of the above two frailty indices in predicting life-threatening morbidity and mortality among older patients following elective high-risk abdominal surgery.This retrospective cohort study extracted perioperative data of older patients (age ≥65 years) undergoing elective high-risk abdominal surgery at a single institution between January 2018 and December 2020. Preoperative frailty was screened by mFI and RAI-rev scoring systems. The primary outcome was the composite of postoperative life-threatening morbidity and mortality during hospitalization. Multivariable logistic regression analyses were performed to investigate the association of the two frailty indices with the primary outcome. Receiver-operating characteristic (ROC) curve was employed to test the predictive performances of the two frailty instruments in predicting the composite primary outcome. The difference between the area under the curves (AUCs) was assessed by DeLong’s test.p < 0.001; RAI-rev: adjusted OR 1.052 per 1-point increase in score, 95% CI 1.018–1.087, p = 0.002) as well as dichotomized frailty measures  were associated with increased odds of the primary outcome separately. ROC curve analysis showed that the discrimination of mFI and RAI-rev scores for the life-threatening morbidity and mortality was poor and comparable .1,132 older patients  were included. Of these, 107 (9.5%) developed postoperative life-threatening morbidity and mortality. In multivariable logistic regression analyses, rising continuous frailty scores (mFI: adjusted OR 1.319 per 0.09-point increase in score, 95% CI 1.151–1.511, High mFI and RAI-rev scores were associated with an increased risk of life-threatening morbidity and mortality in older patients undergoing elective high-risk abdominal surgery. However, both frailty indices displayed poor discrimination for postoperative life-threatening morbidity and mortality. With the rapid expansion of the aging population, frailty has constituted a critical public health issue for healthcare providers worldwide. Frailty is a multidimensional geriatric syndrome characterized by reduced physiologic reserve, accumulated deficits, and decreased resistance to stressors , 2. WithCompared with low-risk surgery, high-risk surgery exerts greater physiologic stress on older individuals and is prone to higher odds of major morbidity and mortality . It is iOver the past decades, dozens of frailty assessment tools have been developed , 14–21. The present study aimed to compare the performances of mFI and RAI-rev in predicting the composite outcome of life-threatening morbidity and mortality in older patients who underwent elective high-risk abdominal surgery.This retrospective cohort study was conducted in Peking University First Hospital, a tertiary general hospital in Beijing, China. The ethical approval was provided by the Biomedical Research Ethics Committee of Peking University First Hospital (2019 [296]). The Ethics Committee agreed to waive the written informed consent from the participants due to the retrospective nature of the study and that no patient follow-up was carried out. The privacy of participants was strictly observed.Older patients (≥65 years of age) who received elective high-risk abdominal surgery  in Peking University First Hospital from January 2018 to December 2020 were screened for study inclusion. We utilized the Operative Stress Score (OSS) system to select patients who underwent high-risk surgery. OSS system rates common operations according to physiologic stress, i.e., OSS1, very low stress; OSS 2, low stress; OSS 3, moderate stress; OSS 4, high stress; and OSS 5, very high stress . We defiThe 11 frailty deficits contained in the mFI were collected based on the National Surgical Quality Improvement Program definitions; each component was allocated the same weight of 1 point . The mFIRAI-rev score was obtained by evaluating 11 variables, i.e., age, sex, cancer, unintentional weight loss, poor appetite, renal failure, congestive heart failure, shortness of breath, residence other than independent living, functional status, and cognitive decline. The weight of each item is detailed in Baseline characteristics not covered by mFI or the RAI-rev were collected, including ASA physical status classification, body mass index (BMI), smoking status, current alcoholism, other major comorbidities, and main laboratory test results. Intraoperative data were also gathered, including risk stratification of surgery categorized by OSS  , duratioThe primary endpoint was the composite postoperative outcome of life-threatening morbidity and mortality during hospitalization, i.e., defined as Clavien-Dindo (CD) grade IV and V complications . CD gradχ2 tests, continuity-corrected χ2 tests, or Fisher’s exact tests. Time-to-event variables were analyzed using the Kaplan–Meier estimator, with differences between groups assessed by the log-rank test.The baseline and perioperative variables were compared between patients with life-threatening morbidity and mortality and those without. We also compared the postoperative outcomes between frail and non-frail patients according to the dichotomized frailty measures. Continuous variables were analyzed with the independent samples t-test or Mann–Whitney U test; the Kolmogorov–Smirnov test was performed to check for normality. Categorical variables were analyzed using Due to the non-normal distribution of the data, we used Spearman’s correlation analysis to test the correlation of the two continuous frailty measures. The agreement of dichotomized measures was evaluated using the percentage of agreement and Cohen’s Kappa coefficient.p values <0.10 in univariate analyses were then included in a multivariable model to examine the covariate-adjusted relationship between the rising mFI score and the primary outcome. Next, we analyzed the mFI score as a dichotomized measure and built another multivariable model to determine the adjusted association of frailty with the primary outcome. Likewise, the above statistical method was employed to explore the relationship of RAI-rev scores with the primary outcome. Herein, we defined each unit of the RAI-rev score as 1 point; similarly, the 11 variables included in RAI-rev were not enrolled in the corresponding multivariable models. All the multivariable analyses were performed with the backward stepwise method.We used univariate and multivariable logistic regression analyses to calculate odds ratios (ORs) and 95% confidence intervals (CIs) of frailty in predicting life-threatening morbidity and mortality. We first analyzed the mFI score as a continuous variable and calculated the ORs and 95% CIs for the primary outcome per one-unit increase in mFI scores. Herein, to facilitate the clinical application of our findings, we defined one unit of the mFI score as 0.09 points . Potential confounding factors (not including the 11 variables covered by mFI) were screened by univariate analyses and tested for multicollinearity by variance inflation factor analysis. Factors with p < 0.10 in log-rank tests) were included in multivariable Cox proportional hazards regression models to examine the adjusted relationship of frailty with the primary outcome. The factors included in the mFI and RAI-rev were not entered into the corresponding multivariable Cox regression model.Besides, we conducted survival analysis to further explore the time effect of frailty  on postoperative life-threatening morbidity and mortality. Herein, we adopted the 30-day life-threatening morbidity and mortality after surgery as the primary outcome in the survival analysis since almost all the primary endpoint events occurred within 30 days postoperatively in our study. The time to the endpoint event was calculated from the time of surgery to the date of the occurrence of life-threatening morbidity and mortality. Patients who were not observed to experience the primary outcome within 30 days and remained hospitalized after 30 days as well as those who did not undergo the endpoint event during hospitalization and were discharged from hospital within 30 days were all censored accordingly. Univariate analyses were performed using the Kaplan–Meier estimator with comparisons between frail and non-frail patients assessed by log-rank test. After multicollinearity screening, potential confounding factors (set at The predictive performances of mFI and RAI-rev were tested using the receiver-operating characteristic (ROC) curve analysis. The area under the ROC curve (AUC) was measured to test the discriminative power (ability to classify correctly) for the primary outcome. An AUC value of 1 indicates the best discrimination, whereas a value of 0.5 indicates that the predictor is no more reliable than chance. Generally, a predictor may be considered useful in clinical decision-making when the AUC exceeds 0.7 . Differep values <0.05 were considered statistically significant. p-values were not corrected since no multiple comparison test was involved. All statistical analyses were performed with the SPSS version 26.0  and MedCalc version 19.05 .For all analyses, two-tailed From January 2018 to December 2020, 2,165 older patients (≥65 years of age) who experienced elective high-risk general and urologic surgical procedures were screened. Of these, 1,033 patients were excluded because they met the exclusion criteria , leaving 1,132 patients in our analysis cohort .The study population had a mean age of 73.4 ± 6.2 years; 63.9% (723/1132) were male. The median mFI and RAI-rev values of our cohort were 0.09 [IQR: 0.09–0.18] and 38 [IQR: 37–43], respectively. The distribution of the mFI and RAI-rev scores across the cohort is displayed in p < 0.05). Similarly, there were significant differences in the above outcomes between the patients with an RAI-rev score of <45 and those with a score of ≥45 ; whereas we found no significant difference in mortality between the two mFI subgroups , indicating a low correlation between the two frailty indices. The overall percentage of agreement between the two dichotomized measures was 72.2% (817/1132). Specifically, 102 (9.0%) and 715 (63.2%) patients were identified as frail and non-frail by both measures, respectively. Cohen’s Kappa test showed slight agreement between the two dichotomized measures  increase in the mFI score  after correcting for the above confounding factors  after adjustment for the above confounding factors  patients, the frail patients (determined by mFI scores of ≥0.27) had a shortened time to develop 30-day life-threatening morbidity and mortality  and 0.613 (95% CI 0.583–0.641), respectively. Although the AUC of RAI-rev was slightly higher than that of mFI, no statistical difference between them was detected .As seen in This retrospective cohort study determined that rising mFI and RAI-rev scores were associated with a higher risk of life-threatening morbidity and mortality in older patients after elective high-risk abdominal surgery. However, both the two frailty indices performed poor discriminative abilities for the occurrence of life-threatening morbidity and mortality.In the present study, we found a low correlation and slight agreement between the two frailty indices. As shown in The effect of frailty on major postoperative morbidity and mortality has been extensively studied , 24, 27.In the present study, ROC analysis results demonstrated that neither the mFI nor the RAI-rev was equipped with good discrimination for serious morbidity and mortality in older patients undergoing elective high-risk abdominal surgery. Our findings are congruent with previous studies that applicated frailty indices to predict postoperative morbidity , 9, 27. Our findings demonstrated clinical significance and might play an important role in perioperative settings. Based on the multivariable logistic and Cox regression analysis results, frailty was significantly associated with a higher risk of life-threatening morbidity and mortality in older patients after elective high-risk abdominal surgery. This finding can help clinicians forecast the elevated risk of serious postoperative outcomes in frail older patients and improve the shared decision-making process. Once a patient is screened as frail, determining the goals of care and selecting the optimal approach to achieve the goals of care constitute crucial components of shared decision-making. It should be carefully considered whether aggressive surgical intervention or palliative care can get frail patients to their goals of care. Realistic expectations and appropriate decision-making may, in turn, decrease perioperative mortality . FurtherBesides the retrospective nature, our study had several notable limitations. First, as mentioned above, the patients with severe frailty were inevitably excluded from the study cohort due to the high-risk nature of the surgery, which might lead to selection bias. Second, the primary outcome was limited to in-hospital serious morbidity and mortality. The complications below CD grade IV were not included in our analysis and the post-discharge outcomes were not gathered, which might underestimate the rate of adverse outcomes. Finally, in our study, the composite endpoint rate was fairly low, especially mortality. Our relatively limited sample size could not enable us to fully elucidate the relationship between frailty and mortality. Despite these, our results demonstrate clinical significance and generate clues for further investigation.In conclusion, this study determined that high mFI and RAI-rev scores were associated with an increased risk of life-threatening morbidity and mortality in older patients following elective high-risk abdominal surgery. However, both frailty indices displayed poor discrimination for the composite outcome of life-threatening morbidity and mortality.The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.The studies involving human participants were reviewed and approved by The Biomedical Research Ethics Committee of Peking University First Hospital. The ethics committee waived the requirement of written informed consent for participation.C-QL contributed to the conception, study design, and supervision. C-QL, HK, and Z-ZX collected data. C-QL, J-HM, and X-YL analyzed and interpreted the data. C-QL drafted the manuscript. All authors critically revised the manuscript, agreed to submit it to the current journal, and gave final approval of the version to be published.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.


Objective: Minimally invasive direct coronary artery bypass grafting (MIDCAB) using the left internal thoracic artery to the left descending artery is a clinical routine in the treatment of coronary artery disease. Far less is known on right-sided MIDCAB (r-MIDCAB) using the right internal thoracic artery (RITA) to the right coronary artery (RCA). We aimed to present our experience in patients with complex coronary artery disease who underwent r-MIDCAB. Materials and Methods: Between October 2019 and January 2023, 11 patients received r-MIDCAB using RITA to RCA bypass in a minimally invasive technique via right anterior minithoracotomy without using a cardiopulmonary bypass. Underlying coronary disease was complex right coronary artery stenosis (n = 7) and anomalous right coronary artery . All procedure-related and outcome data were evaluated prospectively. Results: Successful minimally invasive revascularization was achieved in all patients (n = 11). There were no conversions to sternotomy and no re-explorations for bleeding. Furthermore, no myocardial infarction, no strokes, and, most importantly, no deaths were observed. During the follow-up period (median 24 months), all patients were alive and 90% were completely angina free. Two patients received a repeated revascularization after surgery but independently from the RITA-RCA bypass, which was fully competent in both patients. Conclusion: Right-sided MIDCAB can be performed safely and effectively in patients with expected technically challenging percutaneous coronary intervention of the RCA and in patients with ARCA. Mid-term results showed high freedom from angina in nearly all patients. Further studies with larger patient cohorts and more evidence are needed to provide the best revascularization strategy for patients suffering from isolated complex RCA stenosis and ARCA. Treatment of coronary artery disease (CAD) is being performed using interventional or surgical techniques. Percutaneous coronary intervention (PCI) has become rapidly advanced and broadly available, leading to guideline adoption and an overall reduction in surgical myocardial revascularization. Especially isolated single- and double-vessel CAD is predominantly treated by PCI. Nowadays, MIDCAB has become a well-established surgical routine treatment for proximal left anterior descending (LAD) stenosis, chronic totally occluded LAD, or in high-risk patients with multivessel CAD as an isolated treatment or as part of a hybrid revascularization strategy ,2.Characterized by a small skin incision and access (left anterior minithoracotomy), avoidance of sternotomy and no need for cardiopulmonary bypass could prove favorable, leading to significantly shorter ventilation time, less perioperative trauma, and thus a reduction in hospital length of stay ,4.While left-sided MIDCAB using the left internal thoracic artery (LITA) for grafting the LAD has been widely reported, there is less evidence about right-sided MIDCAB (r-MIDCAB) using the right internal thoracic artery (RITA) for grafting the right coronary artery (RCA). Isolated surgical revascularization of the RCA is very rare and, in most cases, treated by PCI. However, there are still some specific indications in complex coronary artery disease where the surgical approach is the only possible option.The aim of the current study was to evaluate the feasibility and safety, as well as short- and mid-term outcomes of r-MIDCAB in patients with complex CAD of the RCA.This single-center study involves 11 patients who underwent isolated r-MIDCAB procedures at our institution between October 2019 and January 2023. All patients received RITA to RCA bypass using a minimally invasive technique via right anterior minithoracotomy without using a cardiopulmonary bypass. Indications for surgery included the following: 1. complex CAD in seven patients; and 2. anomalous right coronary artery (ARCA) in four patients.All patients received a routine preoperative coronary angiography, transthoracic echocardiography (TTE), and standard chest X-ray to determine the right cardiac silhouette. In addition, patients received cardiac computed tomography (CT) to facilitate pre- and post-operative planning. 3Mensio Structural Heart, version 10.2. (SP1-Build 5233.47)  was used for three-dimensional reconstruction of the CT scans for checking the feasibility of the minimally invasive approach and for optimal planning, including the position of the skin incision and minithoracotomy .In all patients, unilateral lung ventilation, established by double-lumen tracheal intubation or by selective endoluminal blockage under bronchoscopic guidance, was performed. Transesophageal echocardiography was used in all patients. The patients were placed in the supine position on the table, and the chest was elevated 30° on the right side with a soft pillow underneath the scapula.2 blower for visual clearance of the operative field. After completion of the anastomosis, the bypass was checked by transit time flow measurement (TTFM) . The pericardium was closed and a 28F chest drain placed into the right pleural cavity; thereafter, the lung was reinflated under direct vision. The minithoracotomy was closed using a costal resorbable polyglyactin 2-0 Z-suture, followed by the layered wound closure and administration of periosteal and intramuscular 0.7% bupivacaine.Right-sided MIDCAB was performed through a 5–6 cm long right anterior minithoracotomy, and RITA was harvested under direct vision using a Braun/Hauser retractor  . The RITMost patients were extubated in the operating room after checking their neuromuscular state and transferred to the intensive care unit afterwards. Acetylsalicylic acid at a dose of 100 mg per day was started 1 day after surgery and is recommended for life. In addition, dual antiplatelet therapy with either Ticagrelor 90 mg twice daily or Clopidogrel 75 mg daily was started on the second postoperative day for 12 months after surgery to enhance the graft patency rate.Pre-, peri-, and postoperative data were collected prospectively. Follow-up was performed in all patients by out-patient visits or phone calls, and additional information was provided by general physicians and cardiologists. Particularly, data on survival, freedom from angina, number of hospitalizations, and freedom from major cerebral and cardiac events were collected.Continuous and discrete variables were expressed as mean ± SD or median and interquartile range (IQR) for data not normally distributed. Categorical and ordinal variables were expressed by the number and percentage of observations. A statistical analysis was performed using Stat view  and IBM SPSS .Patients’ demographics and preoperative data are summarized in The intra- and early postoperative outcomes are presented in The detailed follow-up data is presented in The right-sided MIDCAB procedure is rarely described in the literature, and we present our rather large experience with this technique in patients with complex CAD. In analogy to the well-established MIDCAB technique with LIMA-LAD grafting, we believe r-MIDCAB is a very valuable technique for minimally invasive treatment of complex CAD of the RCA in selected patients. There were two main indications for the surgery (ARCA and CAD) in our cohort, and we also focused on the perspectives for both.ARCA is a very rare coronary anomaly and has a prevalence of 0.23% in the general population . The potThe data from our study support the safety and feasibility of the r-MIDCAB in ARCA. There was no conversion to sternotomy and no 30-day mortality in our cohort. One patient received acute repeated PCI due to early in-stent thrombosis of the LAD, which was implanted 4 weeks prior to surgery. The RITA-RCA bypass was fully patent and had perfect run-off. During the follow-up, no recurrent angina was observed in three out of four patients, and no further treatments were needed. One patient was re-hospitalized due to polymorphic symptoms and classified as grade CCS II. He underwent a cardiac CT scan followed by catheterization, showing a patent RITA-RCA bypass with excellent run-off . The firThe regular CAD patients in this series were all symptomatic with persistent angina ranging in CCS grade II-III. The majority of the patients (86%) had prior PCI attempts. Interventional treatment was associated with either unsuccessful revascularization due to a demanding anatomy (43%) or led to in-stent re-stenosis  in a dominant RCA. As expected, all of the patients had a relatively low mean SYNTAX score of 8.9 ± 6.0 and were therefore not initially referred to surgical revascularization. The treatable lesions with the r-MIDCAB technique are in RCA segments one and two and, depending on anatomy, in proximal segment three . The patTo the best of our knowledge, the current series is the largest cohort of r-MIDCAB patients that has been published yet.This study has several limitations. It is based on a single-center experience with a new operative technique, the number of patients is very small and the follow-up is just mid-term.This series provides evidence that r-MIDCAB is feasible and safe, with very good mid-term results. More evidence is needed to provide the best revascularization strategy to patients suffering from isolated RCA stenosis and ARCA; therefore, larger and possibly multicenter studies with prospective randomized trials are warranted.