Intermediate endpoints, such as disease-free survival (DFS), have shown good correlation with overall survival (OS) in early-stage non–small cell lung cancer (NSCLC) clinical trials. However, real-world data are limited, and no previous real-world study has quantified the clinical and economic burden of disease recurrence.To examine the association between real-world DFS (rwDFS) and OS and quantify the association between NSCLC recurrence and health care resource utilization (HCRU), health care costs, and OS in patients with resected early-stage NSCLC in the United States.Data from the Surveillance, Epidemiology, and End Results–Medicare database (2007-2019) for patients with newly diagnosed stage IB (tumor size ≥ 4 cm) to IIIA (American Joint Committee on Cancer 7th edition) NSCLC who underwent surgery for primary NSCLC were analyzed in this retrospective observational study. Baseline patient demographic and clinical characteristics were described. rwDFS and OS were compared between patients with vs without recurrence using Kaplan-Meier curves and the log-rank test; their correlation was assessed using normal scores rank correlation. All-cause and NSCLC-related HCRU and health care costs were summarized, and mean monthly allcause and NSCLC-related health care costs were compared between cohorts using generalized linear models.P < 0.001). OS and rwDFS were significantly correlated . Patients with recurrence also had significantly higher all-cause and NSCLC-related HCRU and mean monthly all-cause and NSCLC-related health care costs during the study period.Of the 1,761 patients who underwent surgery, 1,182 (67.1%) had disease recurrence; these patients had shorter OS from the index date and shorter subsequent OS at each postsurgery landmark  than those without recurrence  whose disease comes back are more likely to die sooner. They also use more health care resources and spend more on health care than those who remain tumor-free. The study also showed that patients whose disease comes back later are more likely to die later. These findings highlight the need to prevent or delay NSCLC recurrence to potentially reduce the negative consequences. Implications for managed care pharmacyRecurrence of NSCLC imposes a substantial clinical and economic burden on patients and society. The results of this real-world study highlight a need for more effective treatments that prevent or delay recurrence and prolong survival in patients and alleviate the health care burden of early-stage NSCLC. They also show that disease-free survival is a useful endpoint that can help to optimize managed care pharmacy practice by providing timely data on treatment effectiveness.1 Non–small cell lung cancer (NSCLC) is the most common type of lung cancer, accounting for more than 85% of cases.2Lung cancer is one of the most common and deadly cancers in the United States. The American Cancer Society projects approximately 230,000 lung cancer diagnoses and more than 130,000 deaths from the disease in 2023.4 Patients with resectable stage IB (AJCC 7th edition) disease with tumors 4 cm or larger or other high-risk factors are treated with a similar approach as patients with stage II-IIIA disease.5 The treatment landscape in the adjuvant setting has dramatically evolved with the advent of novel treatments, including targeted therapy and immunotherapy. In the United States, atezolizumab and osimertinib have been approved for adjuvant therapy after resection in patients with stage II to IIIA NSCLC whose tumors have PD-L1 expression on at least 1% of tumor cells in 2021 and in patients with NSCLC whose tumors have epidermal growth factor receptor exon 19 deletions or exon 21 L858R mutations in 2020, respectively.7 Pembrolizumab was approved in 2023 for adjuvant treatment following resection and platinum-based chemotherapy for stage IB (T2a ≥ 4 cm), II, or IIIA NSCLC.8The management of NSCLC varies by disease stage. The standard primary treatment is radical surgery followed by up to 4 cycles of adjuvant therapy with a platinum-based doublet for patients with resectable stage II-IIIA (American Joint Committee on Cancer [AJCC] 7th edition) disease.11 A positive correlation between DFS and OS in early-stage NSCLC has been demonstrated using clinical trial data12; however, there is no such evidence from real-world studies.Given the prolonged timeline and significant costs associated with demonstrating overall survival (OS) benefit in early-stage cancer, intermediate endpoints, such as disease-free survival (DFS), have been used as primary endpoints in clinical trials of adjuvant therapies for early-stage NSCLC.14 there have been few studies on the economic burden and OS associated with recurrence in early-stage NSCLC in a real-world patient population. As the treatment landscape for early-stage NSCLC evolves, such evidence is needed to evaluate clinical and economic impact of novel therapies that can potentially prevent or delay recurrence and improve patient outcomes.Although health care resource utilization (HCRU) and health care costs associated with NSCLC have been documented,This retrospective observational study was conducted using Surveillance, Epidemiology, and End Results (SEER)–Medicare linked data to examine the association between real-world DFS (rwDFS) and OS and quantify the association between NSCLC recurrence and HCRU, health care costs, and OS in patients with resected early-stage NSCLC in the United States.Patients with newly diagnosed stage IB (tumor size ≥ 4 cm)-IIIA (AJCC 7th edition) NSCLC were selected from the SEER-Medicare database (2007-2019). The database links data from 2 sources—namely, SEER cancer registries  and Medicare insurance claims for health care services used by beneficiaries from the time of eligibility until death. As of August 2022, the database included all Medicare-eligible persons in the SEER registry diagnosed with cancer through 2017 and their Medicare claims through 2019. Information on Medicare drug coverage (Part D) was available starting from 2007. All data were limited and adhered to the Health Insurance Portability and Accountability Act and Declaration of Helsinki.16 For patients without recurrence, the index date was a randomly selected date based on the distribution of time between the first surgery and recurrence in patients who experienced a recurrence. The baseline period was the 12 months preceding the index date, and the follow-up period was the time from the index date to the earliest among end of data availability, end of enrollment, and death . Patients had to have undergone their first surgery for NSCLC before 2015 to ensure a theoretical minimum follow-up of 5 years (from date of primary surgery to data cutoff).The study sample consisted of patients who underwent surgery for primary lung cancer with or without adjuvant chemotherapy and who did not receive chemotherapy in the neoadjuvant setting or adjuvant radiotherapy. The index date for patients with recurrence was defined as the date 30 days before the record of recurrence to capture claims in the study period that were likely recurrence-related but occurred before recurrence was observed in the claims data.International Classification of Diseases for Oncology, Third Edition codes 340 - C343, C348, and C349 and relevant histology codes ; (2) diagnosed with stage IB (tumor size ≥ 4 cm)-IIIA NSCLC based on AJCC 7th edition tumor node metastasis staging criteria17 , aged 66 years or older at the time of diagnosis, and with continuous Medicare Parts A, B, and D enrollment for at least 12 months afterward; (3) no other cancers before the diagnosis of NSCLC, continuously enrolled in Medicare Parts A, B, and D for at least 12 months after the initial diagnosis, and underwent lobectomy, bilobectomy, or pneumonectomy within 6 months of NSCLC diagnosis; (4) had not previously received neoadjuvant chemotherapy and/or neoadjuvant or adjuvant radiotherapy; and (5) had no diagnoses of distant secondary malignancies before or within 30 days after surgery. Patients with health maintenance organization coverage anytime during the study period were excluded.The flow diagram of patient selection is shown in Patients who met the inclusion criteria were assigned to the recurrence or nonrecurrence cohort. Recurrence after surgery  was defined as a diagnosis of metastatic disease, new diagnosis of locoregional disease, and/or any additional treatments for NSCLC  following a 90-day treatment-free interval starting immediately after the initial surgery. Adjuvant chemotherapy, which was defined as any US Food and Drug Administration–approved or National Comprehensive Cancer Network–recommended adjuvant treatment initiated within 90 days after the surgery, was permitted during the 90 - day treatment-free interval. For patients who received adjuvant treatment, the following 2 events were further considered to identify recurrence: (1) initiation of a new regimen at least 30 days after the start of adjuvant treatment; (2) reinitiation of the same regimens at least 90 days after the end of adjuvant treatment.Baseline patient demographic and clinical characteristics were collected. Clinical outcomes were rwDFS and OS. rwDFS was defined as the time from the initial surgery for NSCLC to first recurrence or death, whichever occurred first. For analyses of the burden associated with disease recurrence, OS was defined as the time from the index date to death. For the analysis of the correlation between OS and rwDFS, OS was defined as the time from the initial surgery to death. In the landmark analysis assessing the association between OS and rwDFS, OS was defined as the time from a predetermined landmark time point  to death. For all outcomes, patients were censored at the earlier of the last date of follow-up in the SEER-Medicare data and end of data availability.18All-cause NSCLC-related HCRU and health care costs were determined based on claims for medical services or for the administration of antineoplastic treatments associated with a diagnosis for NSCLC. Specifically, a claim was considered NSCLC-related if there was a primary diagnosis of NSCLC for an outpatient visit. For other types of visit, a primary or a secondary diagnosis of NSCLC was used to determine whether the claim was NSCLC-related. All-cause and NSCLC-related HCRU outcomes were summarized as incidence rate (IR) per patient per month (PPPM), defined as the ratio between total number of events and total person-months during the follow-up period to account for variable lengths of follow-up in the study period across individual patients. NSCLC-related costs were costs associated with NSCLC-related HCRU, including medical costs and NSCLC-related Part D pharmacy costs. All costs were inflated to 2021 US dollars using the personal consumption expenditures medical care component.P values. Mean monthly all-cause and NSCLC-related health care costs per patient were compared between patients with and without recurrence using generalized linear models with a Tweedie distribution and log-link function. For both HCRU and health care costs, the covariates were age at index date, sex, race, region, CCI, histology, disease stage at diagnosis (AJCC 7th edition staging), and number of all-cause inpatient and outpatient visits PPPM during the baseline period.Baseline characteristics were described and compared between patients with and without recurrence. Means and SDs are reported for continuous variables, and frequency counts and percentages are reported for categorical variables. Statistical comparisons between the 2 cohorts were performed with the t-test for continuous variables and the chi-square test for categorical variables. OS was described using Kaplan-Meier curves and was compared between the recurrence and nonrecurrence cohorts with the log-rank test. Cox proportional hazards models adjusted for key baseline characteristics, including disease stage at diagnosis (AJCC 7th edition staging), histologic type, age at surgery, sex, race, Charlson Comorbidity Index (CCI), and use of adjuvant chemotherapy. Landmark analyses of OS were performed at 3 time points . Subsequent OS was described and compared between patients with and those without recurrence at landmark time points using Kaplan-Meier curves and Cox proportional hazards models adjusted for key baseline characteristics. The correlation between rwDFS and OS was assessed by normal scores rank correlation. PPPM IRs of all-cause and NSCLC-related HCRU were compared between patients with and without recurrence using generalized linear models with a negative binomial distribution and log-link function. Offset terms were included in the models to account for different follow-up durations. IR ratios (IRRs) comparing outcome measures between the 2 cohorts were reported with corresponding 95% CIs and Overall, 1,761 patients with early-stage NSCLC who underwent surgical resection of the primary tumor, including 1,182 (67.1%) patients with disease recurrence, met the study inclusion criteria . The medP < 0.001). More than half of the patients in both cohorts had nonsquamous disease at initial diagnosis. The proportion of patients who received adjuvant chemotherapy was larger in the recurrence cohort . Patients with recurrence were also younger at the time of surgery  and had a higher CCI  than those without recurrence. In terms of baseline HCRU, patients in the recurrence cohort had a higher number of all-cause outpatient visits  but comparable number of any inpatient visits vs those in the nonrecurrence cohort.The mean age at the index date was approximately 75 years; around half of the patients were male, and 84% in both cohorts were White . A smallP < 0.001) (P < 0.001).The OS from the index date was significantly shorter for patients with recurrence than for those without recurrence  , and theSupplementary Figure 2). At the 1 -, 3 -, and 5 -year landmarks, the median OS was 29.3, 42.3, and 43.4 months, respectively, for patients with recurrence and 77.9, 79.6, and 72.0 months, respectively, for patients without recurrence  (P < 0.001). The estimated normal scores rank correlation of 0.58 indicated that there was a statistically significant correlation between rwDFS and OS .At each postsurgery landmark time point, patients with recurrence had significantly shorter subsequent OS than those without recurrence (< 0.001) . CompareP < 0.001), outpatient visits , emergency department [ED] visits , and days hospitalized . The greatest differences in NSCLC-related HCRU between cohorts were for inpatient admissions , inpatient days , ED visits , and outpatient visits . Accordingly, patients in the recurrence cohort incurred significantly higher health care costs than those in the nonrecurrence cohort (P < 0.001); this implies a potential reduction over 5 years of up to $252,240 per patient in all-cause health care costs, assuming a 5-year delay in disease recurrence.Compared with patients without recurrence, those in the recurrence cohort had significantly higher HCRU during the study period . Specifie cohort . In part19; and in a study of 128 patients with stage I - III NSCLC who underwent complete resection between 2007 and 2018 and experienced disease recurrence, the median postrecurrence survival was 22.5 months.20This retrospective observational study of patients with resected early-stage stage IB (tumor size ≥ 4 cm)-IIIA NSCLC in the United States provides real-world evidence that patients with recurrence have a significantly higher risk of death than those without recurrence. These findings are consistent with previous studies assessing real-world postrecurrence survival among patients with resected early-stage NSCLC. For instance, in a retrospective analysis of 281 patients with stage I-III NSCLC who underwent complete resection between 2005 and 2013 and had recurrent disease, the median postrecurrence survival was 26.1 months12We also assessed the correlation between rwDFS and OS following surgery in patients with early-stage NSCLC. To the best of our knowledge, this is the first study to analyze this correlation using a real-world dataset. The results showed that at each postsurgery landmark time point, patients with recurrence had significantly shorter subsequent OS than those without recurrence and that DFS and OS were positively correlated. The findings are consistent with those of previous analyses of clinical trial data demonstrating a correlation between surrogate survival endpoints and OS in patients with operable and locally advanced lung cancer. For example, a study evaluating DFS, progression-free survival, and locoregional control as potential surrogate endpoints using data for 15,071 patients enrolled in 60 randomized clinical trials concluded, based on a high level of evidence, that DFS is a valid surrogate endpoint for OS in studies of adjuvant chemotherapy involving patients with NSCLC.14; a population-based study comparing HCRU among patients with completely resected stage II-IIIB NSCLC who received adjuvant therapy between 2008 to 2017 at US Oncology Network clinics found that patients had more ED visits and hospitalizations after as compared with before recurrence.21 However, there is limited information on the costs associated with NSCLC recurrence in patients with resected early-stage NSCLC. To address this gap, the present work compared HCRU and health care costs between patients with resected early-stage NSCLC with vs without recurrence. Recurrence was associated with significantly higher all-cause and NSCLC-related HCRU and costs during the study period. This implies that delaying disease recurrence by 5 years with appropriate and effective interventions can yield a saving of up to $252,240 per patient in all-cause health care costs.The economic burden associated with advanced NSCLC has been investigated in other studiesThis study had important strengths. First, the linked SEER-Medicare database provided detailed clinical and demographic information and HCRU for a defined population of patients with resected early-stage NSCLC in the United States. Second, the comparison of OS provided a comprehensive view of the disease burden of patients with vs without recurrence. Third, the demonstration of a correlation between postsurgery rwDFS and OS in patients with early-stage NSCLC further supports the use of DFS to predict overall survival in future studies.22 patients in the SEER-Medicare database are likely to be representative of the general NSCLC patient population in the United States. Patients who did not receive treatment in the event of recurrence or did not have codes associated with metastasis in their claims were also missed. Second, the data incorporated in the analyses spanned from 2007 to 2019, which preceded the approval of any targeted therapy or immunotherapy for NSCLC in the United States; thus, the findings regarding survival outcomes and costs cannot be applied to patients with NSCLC who received these interventions. Third, NSCLC recurrence could not be identified directly from administrative claims data, and an algorithm based on procedure, diagnosis, and drug codes and various assumptions was used for this purpose. As such, coding inaccuracies may have led to misclassification bias and misidentification of patients with NSCLC recurrence. Finally, some clinical information, such as complete resection status, is not available in the data. Therefore, a 90-day treatmentfree interval was used to indicate the end of primary treatment and served as a proxy for the disease-free state.It is also important to acknowledge the limitations to this study. First, the linked SEER-Medicare database only included Medicare beneficiaries aged 65 years or older; therefore, the results may not be generalizable to a younger population of patients with NSCLC. On the other hand, as more than 70% of patients with lung cancer in the United States diagnosed between 2015 and 2019 were aged 65 years or older (with a median age at diagnosis of 71 years),This retrospective, US population-based study found that patients with resected early-stage NSCLC who experienced recurrence had a 3.72 times higher risk of death than those without recurrence . Recurrence was associated with significantly higher all-cause and NSCLC-related HCRU and costs. Additionally, postsurgery rwDFS and OS of patients with resected early-stage resected NSCLC were found to be significantly and positively correlated. Taken together, these findings highlight the importance of preventing or delaying recurrence following surgical resection of early-stage NSCLC to alleviate the clinical and economic burden on patients and the health care system.


Lung cancer is the main cause of cancer mortality worldwide and in Spain. Several previous studies have documented socio-economic inequalities in lung cancer mortality but these have focused on specific provinces or cities. The goal of this study was to describe lung cancer mortality in Spain by sex as a function of socio-economic deprivation.We analysed all registered deaths from lung cancer during the period 2011–2017 in Spain. Mortality data was obtained from the National Institute of Statistics, and socio-economic level was measured with the small-area deprivation index developed by the Spanish Society of Epidemiology, with the census tract of residence at the time of death as the unit of analysis. We computed crude and age-standardized rates per 100,000 inhabitants by sex, deprivation quintile, and type of municipality  considering the 2013 European standard population (ASR-E). We further calculated ASR-E ratios between the most deprived (Q5) and the least deprived (Q1) areas and mapped census tract smoothed standardized lung cancer mortality ratios by sex.We observed 148,425 lung cancer deaths (80.7% in men), with 73.5 deaths per 100,000 men and 17.1 deaths per 100,000 women. Deaths from lung cancer in men were five times more frequent than in women (ASR-E ratio = 5.3). Women residing in the least deprived areas had higher mortality from lung cancer (ASR-E = 22.2), compared to women residing in the most deprived areas (ASR-E = 13.2), with a clear gradient among the quintiles of deprivation. For men, this pattern was reversed, with the highest mortality occurring in areas of lower socio-economic level (ASR-E = 99.0 in Q5 vs. ASR-E = 86.6 in Q1). These socio-economic inequalities remained fairly stable over time and across urban and rural areas.Socio-economic status is strongly related to lung cancer mortality, showing opposite patterns in men and women, such that mortality is highest in women residing in the least deprived areas and men residing in the most deprived areas. Systematic surveillance of lung cancer mortality by socio-economic status may facilitate the assessment of public health interventions aimed at mitigating cancer inequalities in Spain.The online version contains supplementary material available at 10.1186/s12939-023-01970-y. Lung cancer was the main cause of cancer mortality during the year 2020, with an estimated burden of 1.2 million of deaths worldwide [In Spain, lung cancer is the most deadly cancer in both sexes , with 22,930 deaths in 2020 , lung caReducing inequalities in cancer outcomes is a main priority of the European Union (EU) . The EU In Europe, research has documented pervasive socio-economic inequalities in lung cancer mortality using either individual-level  or aTo the best of our knowledge, in Spain, socio-economic inequalities in lung cancer mortality have only been investigated in specific provinces  or withHowever, no study has examined lung cancer mortality in Spain nation-wide and using a comprehensive, small-area index of socio-economic deprivation. Such a study would provide valuable information on the existing inequalities in a geographically diverse area. The recently developed Spanish Deprivation Index (SDI) allows to characterize socio-economic deprivation in Spain and study its impact on diverse health outcomes , 27, 28.All registered deaths from lung cancer during the period 2011–2017 in Spain were included. Lung cancer mortality data (ICD-10 codes C33-C34 ) in SpaiWe used the layer of census tracts of the Spanish Census in 2011. Population data by census tract level, age group, and sex at 1st of January of 2011 were extracted from the National Statistics Institute . PopulatTo measure SES, we used an area-based deprivation index developed by the Spanish Society of Epidemiology, the Spanish Deprivation Index . The SpaThe study protocol has received approval from the Internal Review Board of the Andalusian School of Public Health (CP17/00206), the Granada Provincial Research Review Committee, and the Biomedical Ethics Committee of the Department of Health of the Andalusian Regional Government (study 0072-N-18). The research is in accordance with the principles embodied in the Declaration of Helsinki.Expected lung cancer deaths for each census tract, sex, and age group were computed by multiplying the census tract population of each sex and age group by the overall age-sex-specific mortality crude rate. For each census tract, expected lung cancer deaths of each age group and sex were added to obtain expected lung cancer deaths by sex. For each sex, census tract smoothed standardized lung cancer mortality ratios (SMR) were computed using a conditional autoregressive model  based onObserved crude and age-standardized rates were computed for each quintile of deprivation, considering the 2013 European standard population (ASR-E) . MortaliWe used R v.4.2.0  for dataDuring the period 2011–2017, 151,182 deaths from lung cancer were registered in Spain. Of all these deaths, 1.8% did not merge with the census tract layer and population of 2011, leaving 148,425 lung cancer deaths (80.7% in men) after the merging. Supplementary Fig. The spatial distribution by census tract of the Spanish Deprivation Index quintiles is shown in Fig. Overall SMR by census tract for Spain, without considering SES, is shown in Fig. Table Women residing in the most affluent areas had higher mortality from lung cancer Table , compareThe consistency of the socio-economic inequalities found was supported by a sensitivity analysis using percentiles instead of quintiles to add more detail , differThe mapping of the SMR in our study shows a very similar pattern as other recent work for Spain and Portugal , with thThe socio-economic inequalities in lung cancer mortality remained fairly stable over time and across urban and rural areas. The higher lung cancer mortality we found in urban areas is in accordance with a previous research conducted in the Madrid Autonomous Region , where aThe overall lung cancer mortality rates of our study were aligned with the information provided for Spain in the Global Cancer Observatory (GCO)  and the One limitation of this study is that the deprivation index and population data are referred to 2011, whereas the mortality data covers a longer period (2011–2017). Deprivation may have changed during 2011–2017, nevertheless we argue that aggregated measures of socio-economic inequalities are likely to be fairly consistent over time. Moreover, the use of the European Index of Deprivation  could haWhereas the deprivation index is not based on direct information about income, in a previous study we discovered that, at the census tract level, this deprivation index and the average income per person are correlated in Spain . An addiOne of the main strengths of the study is the use of exhaustive mortality data collected by an official institution. Although some issues with the coding of the cause of death are plausible , 58, forMore broadly, the current results show the need for interventions targeting the social determinants of health, accompanied by regular monitoring over time using standardized measures to better understand the effects of policies on lung cancer mortality. Forthcoming interventions need to consider SES as one of the main components to maximize the effect produced. For instance, Europe’s Beating Cancer Plan promotes the implementation of lung cancer screening with low-dose computed tomography in the European Union . In SpaiA complex approach involving both individual-level and population-level interventions is needed to reduce the extent of the socio-economic inequalities we found in our study. Improving access to healthcare in lower socio-economic groups  and reducing the exposure to risk factors like smoking or environmental pollutants could reduce the inequalities in lung cancer mortality, while more knowledge about the root causes is required for such interventions to be more effective.Socio-economic status is strongly related to lung cancer mortality, showing opposite patterns in men and women, such that mortality is highest in women residing in the least deprived areas and men residing in the most deprived areas. Documenting socio-economic inequalities in lung cancer mortality could help direct policies and interventions aiming to reduce inequalities. Systematic surveillance of lung cancer mortality by socio-economic status may facilitate the assessment of public health interventions aimed at mitigating cancer health inequalities in Spain.Below is the link to the electronic supplementary material.Supplementary Material 1Supplementary Material 2


Glioblastoma (GBM) is a lethal disease. At least in part, the recurrence of GBM is caused by cancer stem cells (CSCs), which are resistant to chemotherapy. Personalized anticancer therapy against CSCs can improve treatment outcomes. We present a prospective cohort study of 40 real-world unmethylated Methyl-guanine-methyl-transferase-promoter GBM patients treated utilizing a CSC chemotherapeutics assay-guided report (ChemoID).Eligible patients who underwent surgical resection for recurrent GBM were included in the study. Most effective chemotherapy treatments were chosen based on the ChemoID assay report from a panel of FDA-approved chemotherapies. A retrospective chart review was conducted to determine OS, progression-free survival, and the cost of healthcare costs. The median age of our patient cohort was 53 years (24–76).P = .011, compared to patients who could be treated with low-response drugs who had instead an OS of 12.5 months (3.0–27.4 months). Patients with recurrent poor-prognosis GBM treated with high-response therapy had a 63% probability to survive at 12 months, compared to 27% of patients who were treated with low-response CSC drugs. We also found that patients treated with high-response drugs on average had an incremental cost-effectiveness ratio (ICER) of $48,893 per life-year saved compared to $53,109 of patients who were treated with low-response CSC drugs.Patients treated prospectively with high-response ChemoID-directed therapy, had a median overall survival (OS) of 22.4 months (12.0–38.4) with a log-rank The results presented here suggest that the ChemoID Assay can be used to individualize chemotherapy choices to improve poor-prognosis recurrent GBM patient survival and to decrease the healthcare cost that impacts these patients. This study shows that ChemoID-predicted chemotherapies that target cancer stem cells and the bulk of tumor cells improve survival outcomes of GBM patients.ChemoID is a functional precision medicine assay that improves the quality of care while reducing healthcare costs.The ChemoID CSC assay improves the outcome of recurrent glioblastoma patients by guiding their chemotherapy treatment and diminishes the cost of healthcare. The ChemoID CSC assay is beneficial in personalizing treatment strategies to increase survival time for recurrent GBM patients and to provide quality metrics for healthcare payers and providers to support access to care.1 Despite this treatment, recurrence is almost inevitable,2 and the prognosis of recurrent GBM remains poor with a median PFS of 5.5 months, and a median OS of 8–9 months.3Glioblastoma (GBM) is the most aggressive brain tumor in adults, exhibiting a very poor prognosis with a median time to recurrence of approximately 7 months, and a median survival of 15–18 months even if treated with standard of care (SOC) consisting of maximal surgical resection, and concurrent radiation with Temozolomide and adjuvant treatment with Temozolomide (TMZ).4–7 In these patients, treatment options depend on specific aspects of its presentation, including secondary cytoreductive surgery when possible, focused re-irradiation,8 and numerous second-line chemotherapy treatment options.9 While most patients eventually succumb to the progression of recurrent disease, second-line chemotherapy treatments have provided variable remission and symptom-free survival in a percentage of patients.9 As such, the selection of effective chemotherapy is extremely important for these patients. Additionally, with emerging value-based payment models where outcomes-based contracts link payment for indications of specific cancer drug prices, there are further concerns about the accessibility and affordability of treatments for recurrent GBM patients; therefore, there is a need for effective anticancer drugs that limit overall cost while also increasing treatment value for these patients.Unfortunately, no universally held SOC is available for recurrent GBM, especially those displaying an unmethylated O6-Methyl-guanine-methyl-transferase (MGMT) promoter and wild-type IDH-1/2 gene status. The methylation status of the promoter region of the MGMT gene along with the status of the IDH-1/2 gene have been indicated as the most important negative prognostic predictors of patients’ outcomes.Herein we describe our experience in using ChemoID, a clinical laboratory improvements amendment (CLIA)-certified and College of American Pathologists-accredited clinical diagnostic test that is performed in a clinical-pathology laboratory, which identifies chemotherapeutic agent(s) that kill both the cancer stem cells (CSCs) and the bulk tumor cells, to guide treatment of recurrent GBM.10 which takes into account parameters and evaluations to identify pseudo-progression as well as pseudo-response. Tumor assessments were performed by an independent neuro-radiology service composed of 2 readers and a third senior reader for adjudication of disagreements. All neuro-radiologists were blinded to groups and/or treatment assignments throughout the study to determine the earliest time of progression independent of the impressions of the treating physicians to avoid bias.Forty patients , 18 years and older, that were clinically diagnosed with unmethylated MGMT-promoter and IDH-1/2 wild-type recurrent GBM received 2 concurrent biopsies, one that was examined by frozen sectioning for histological diagnosis and another for prospective ChemoID chemotherapeutic testing between January 2017 and February 2020 all at the Allegheny Health Network Hospital in Pittsburgh, Pennsylvania. This chart review of prospectively treated patients was approved by the Allegheny Health Network Institutional Review Board. Under a physician’s order, patients were treated with ChemoID-guided chemotherapy according to their overall functional status and ability to tolerate the recommended treatment. Radiological data was collected before surgery, immediately post-surgery, and following chemo/radiation therapy with an MRI follow-up every 2 months. Supportive care was also allowed at the discretion of the treating physician. Disease status was measured by radiologic examination (MRI scan as the primary imaging method), physical examination, and measurements using the Response Assessment in Neuro-Oncology (RANO) criteria,11–15 In brief, recurrent GBM participants underwent surgical resection of the recurrent tumor and fresh tissue biopsy samples were collected in the operating room under sterile conditions and divided into 2 parts. One part of the biopsy was sent overnight via FedEx clinical pack in a sterile vial containing a transportation medium to the clinical-pathology laboratory at Cabell Huntington Hospital in West Virginia under physician order to perform the ChemoID assay under Clinical Laboratory Improvement Amendments (CLIA)-certified and College of American Pathologists. The second portion of the biopsy was placed in a 10% formaldehyde solution and sent to the local pathology lab for histopathological confirmation. Tissue samples from recurrent GBM-confirmed tumors were also evaluated for methylation of the MGMT gene promoter. Post-surgery/biopsy, patients received a baseline contrast-enhanced brain MRI or CT if MRI was contraindicated.Details regarding the CSCs cytotoxicity assay (ChemoID) procedure have been described previously.2 were kept consistently constant in an incubator. The bioreactor can rotate at adjustable speed on a fixed axis creating a 3D-suspension cell culture in the absence of shear forces. Primary cells were counted and 2 × 10^6 cells were cultured in the bioreactor for 7-day set at 25 rpm with airflow set at 20% in RPMI media in the absence of growth factors.15 Plates (96-well) were seeded with equal numbers of either bulk tumor cells or CSCs and incubated at 37°C. After 24 hours, clinical-grade chemotherapy drugs were added alone or in combination for 1-hour exposure (Table 1). After the 1-hour exposure, the treatment media containing the various chemotherapies were removed and replaced with fresh media. Cell viability was assessed 48 hours later as previously.15 For each treatment, percent survival  was calculated relative to appropriate controls. Efficacy and resistance of each drug and combinations were reported on the ChemoID assay results as a continuous number from <10% to 100% cell kill.15To generate the primary tumor cell cultures the fresh tumor tissue from surgical biopsies was minced and gently disassociated in a biosafety cabinet. The CSCs were enriched from the primary tumor cell cultures using a 3D-suspension cell culture rotating bioreactor with a volume of 40 mL and a gas-permeable membrane that allows for gas exchange. Culture media, oxygenation, rotation speed, temperature, and CODescriptive statistics were constructed as medians with ranges for continuous variables and counts and percentages for categorical variables.Two different responder categories were defined: The bulk of tumor responders were those subjects who received a treatment identified by the drug response assay as 55% or above cell kill for the bulk of the tumor and CSC responders were those subjects who received a drug in which the test identified as 40% or above cell kill of CSCs. These cell kill values were derived from previous research and validated in this sample via Youden indices. Summary statistics were calculated where appropriate and all relevant graphs were constructed. Kaplan–Meier graphs were constructed and hazard ratios were calculated. Model assumptions were graphically checked and tested via Schoenfeld residuals and were found to be satisfactory. All statistical analyses were completed using Stata v15.1 .The major costs associated with the treatment of recurrent GBM are the cost of the drug response assay, surgery, chemotherapy, adverse events and toxicities, and of end-of-life care. It was assumed that all our patients incurred the same cost per surgery, end-of-life care, and treatment of toxicities; therefore, we did not include these costs in our analysis. For our analysis, we considered only the cost of chemotherapies and the cost of hospitalization due to chemotherapy adverse events. The main source for the cost data in this analysis reflects current Medicare pricing.17All patients in the current model were prospectively treated with a chemotherapy regimen appropriate to their GBM at the time of their recurrence. The costs associated with 6 cycles of each chemotherapy regimen, as well as the associated administration costs (in the physician office setting), were estimated using the current Medicare physician fee schedule for administration payments and drug pricing database for chemotherapy agents.18 Keeping the costs for surgery, ChemoID assay, end-of-life care, and adverse event treatment constant between our cohort and historical data, the model results are affected only by the cost of chemotherapies and by survival outcomes. To account for the uncertainty in the hazard ratio estimates associated with the assay and its impact on the ICER/LYS, the range in the ICER/LYS was estimated by 1000 bootstrap samples for the assay. Several stratified analyses for the reference model are also reported. To assess the sensitivity of the model due to the cost of chemotherapy, the scenario when the oncologist chooses the least expensive treatment within the highest category of sensitivity for each patient in the assay-consistent cohort was also investigated.The relative cost-effectiveness of the intervention is expressed by the incremental cost-effectiveness ratio per life-year saved (ICER/LYS), which is the ratio of the difference in the average costs per patient to the difference in the mean overall survivals. The standard threshold for a healthcare intervention to be deemed cost-effective is an expenditure of between $50,000 and $100,000 per additional year of life saved.4–7 The ChemoID assay was performed on all patients using a validated panel of the chemotherapies listed in P = .277).Fresh tissue samples were collected for ChemoID testing from 40 patients affected by unmethylated-MGMT-promoter GBM, with an overall median age of 53 years (range 24–76), 77% of which were male, all eligible for surgical resection. Recurrence of GBM was confirmed on frozen sections for all patients. The methylation status of the promoter region of the O6-MGMT gene was studied for our entire patient cohort. All patients analyzed in this study had an unmethylated MGMT gene promoter status, indicating they had a negative prognostic predictor of outcome.P = .059). Median tumor volume at recurrence was 13.6 cm3 for the low-response treatment group and 16.1 cm3 for the high-response treatment group (P = .210). The median amount of steroid prescription for a 24-hour period was 4 mg for the low-response treatment group and 2 mg for the high-response treatment group (P = .640).The overall median KPS status of our patients was 70 (KPS range 60–90). Median KPS was 90 for the group of patients treated with low-response CSC drugs and 80 for the high-response treatment group (P = .627).Next-generation DNA sequencing was conducted on samples from all patients in our cohort. In 80% of the patients in the low-response treatment group and 82% of the patients in the high-response group, the next-generation DNA sequencing revealed an actionable drug . All patients have been followed up by MRI every 2 months, and the median follow-up from tumor biopsy time was 15.8 months (range 2.0–56.0 months). At the end of our follow-up, 11 patients were still alive. P = .011. Patients with recurrent GBM treated with high-response therapy had a hazard of death of 0.37 with a 95%CI (0.16–0.82), P = .015, compared to patients who were treated with low-response CSC drugs. Notably, of the surviving patients, all have exceeded the expected survivals of previously reported studies.20We observed that the median OS of patients treated with high-response CSC therapy was 22.4 months (12.0–38.4) compared to patients who were treated with low-response CSC drugs with an OS of 12.5 months (3.0–27.4 months), with a log-rank P = .056. The median time from original diagnosis to recurrence was 11 months for patients who were treated with low-response CSC drugs versus 12 months for high-response drugs (P = .172). Patients with recurrent GBM treated with high-response therapy had a hazard of progression of 0.49 with a 95% CI (0.23–1.06), P = .069, compared to patients who were treated with low-response CSC drugs .Patients were assessed for progression-free survival (PFS) by the RANO. We observed that the median PFS of patients treated with high-response CSC therapy was 14 months (9.0–25.0 months) compared to patients who were treated with low-response CSC drugs with a PFS of 9 months (2.0–22.0 months), with a log-rank 21To better understand the healthcare benefit and the economic impact of the use of the ChemoID assay, we compared the health benefit observed and the cost of therapies used in our patients’ cohort to the historical data of patients treated empirically with chemotherapies, in a similar manner to previously published investigations.Table 3). The mean cost of ChemoID-guided high cell kill anti-CSC therapy administered was $99,221 with a median OS of 22.4 months, average life years saved of 0.55, and an ICER/LYS of $48,893. Instead, the mean cost of low cell kill therapy administered was $57,725 with a median OS of 12.5 months, average life years saved of −0.275, and an ICER/LYS of $57,109. The P-value for the difference in ICER/LYS reported in Table 4). With a mean cost of ChemoID-guided high cell kill anti-CSC therapy administered was $99,221 with a median OS of 22.4 months, the average life years saved was 0.5083, and the ICER/LYS was $72,037. Instead, the mean cost of low cell kill therapy administered was $57,725 with a median OS of 12.5 months, average life years saved of −0.3167, and an ICER/LYS of $15,401.The mean cost of therapies administered to GBM patients in the United States alone is $87,810 with a median OS of 18 months, average life years saved of 0.1833, and an ICER/LYS of $84,436 , followed by radiotherapy with concomitant and maintenance therapy with Temozolomide (TMZ) and/or other secondary chemotherapies,24 Currently, there is insufficient evidence to indicate a superior agent or treatment strategy for recurrent GBM patients as a whole as well as for individual patients.TMZ is a key component of SOC for both newly diagnosed and recurrent GBM patients; however, the major challenge with recurrent GBM treatment is the numerous clinically acceptable and oftentimes equivalent treatment options identified in treatment guidelines.25 The current study evaluated the clinical advantage of using the ChemoID chemotherapeutics assay to measure CSC response against a panel of FDA-approved chemotherapies to treat recurrent GBM. Patients were treated with chemotherapies chosen from those drugs showing the highest cell kill as determined by the ChemoID assay, and by taking into consideration the patient’s tolerability to the indicated treatment. Unfortunately, 11 patients could be treated with low-response drugs against CSCs, and their survival and PFS were decreased when compared to those patients who received high-response CSCs therapy.The presence of CSCs in GBM appears to be responsible, at least in part, for resistance to standard treatments and the variable responses seen to treatment and thus has important implications for the development of a diagnostic assay to guide personalized treatment regimens.4–6 and we observed that patients who were treated with high-response CSC therapy survived 9.9 months longer in median than those patients who could be only treated with low-response CSC predicted drugs.The vast majority of high-response cell kill was seen in this patient cohort when cells were exposed to BCNU-containing regimens . Althoug27 All of the GBM patients in our cohort, received the diagnosis of recurrent disease by histological analysis of a frozen biopsy before providing a sample for the ChemoID assay.Interestingly, systematic reviews and meta-analyses of re-resection and re-irradiation for recurrent GBM indicate that both practices are associated with better overall survival and post-progression survival, providing encouraging disease control and survival rates.29 Interventions such as re-resection, systemic therapy, and/or re-irradiation may benefit selected patients, but unfortunately, all are given with a palliative intent.29 As such, treatment decisions must be individualized. Clinicians are tasked with selecting the most appropriate treatment, balancing the benefits of treatment with the risk of treatment-related toxicity and its impact on the quality of life.31 The performance status of the patient, extent of recurrence , and location of recurrence are important considerations in such instances.31It is known that recurrent GBM is associated with a median overall survival of less than a year and the majority of patients have profound tumor-related symptoms.20 demonstrating the importance of determining CSCs response to chemotherapy to prolong patients’ survival. The data further supports the belief that long-term tumor response in GBM, is in fact, more dependent on the intrinsic sensitivity or resistance of the CSCs to conventional chemotherapies. This concept is especially valuable and important with emerging value-based healthcare models where outcomes-based contracts linked to payment for an indication of specific anticancer-drug prices raise concerns about the accessibility and affordability of treatment for recurrent GBM patients. In the healthcare benefit analysis of this cohort of MGMT unmethylated recurrent GBM patients, the majority of ChemoID assay high-response chemotherapies were observed when cells were exposed to regimens containing BCNU, one of the least expensive drugs on the list of those considered. The analysis would have resulted in a completely different conclusion if, in turn, the most expensive drugs were also found to be the most effective. Nevertheless, our data showed that BCNU-containing regimens, as predicted by the ChemoID assay, were clinically advantageous in this cohort of MGMT unmethylated recurrent GBMs. The power of precision medicine lies in its ability to guide healthcare decisions toward the most effective treatment for a given patient, thereby improving healthcare quality while reducing the need for unnecessary therapies and lowering costs.Notably, our cohort of poor-prognosis GBM patients treated with assay-guided therapy had a 63% probability of survival at 12 months, compared to the 27% historical probability of survival at 12 months observed in previous studies,12 Targeting of CSCs alongside the bulk of other cancer cells is a new paradigm in personalized anticancer treatment. This strategy and technological advancement constitute an important advantage of the ChemoID approach over other diagnostic methods for personalized medicine since individual chemotherapies or their combinations are functionally tested on the patient’s cancer cells and CSCs.ChemoID is a functional precision medicine test that uses a patient’s live bulk of tumor cells and CSCs isolated by tumor biopsies to determine which chemotherapy agent (or “combinations”) is most effective.32–35 The study was designed as a parallel-group, controlled clinical trial that randomized participants to either standard-of-care chemotherapy chosen by the physician or ChemoID-guided therapy. In the randomized clinical trial, response to therapy was measured by MRI imaging using RANO criteria to assess overall survival (OS), OS at 6, 9, and 12 months, median PFS, PFS at 4, 6, 9, and 12 months, objective tumor response, time to recurrence, and quality of life. In the randomized study, the recurrent GBM patients treated with ChemoID-guided therapy survived 4.5 months longer in median than those treated with chemotherapies empirically chosen by the physicians.We have also conducted a multi-institutional randomized clinical trial (NCT03632135) and determined the clinical validity of the ChemoID assay as a predictor of clinical response in recurrent GBM.ChemoID is the first and only chemotherapeutics assay currently available in oncology clinics that examines CSCs susceptibility to conventional FDA-approved drugs from solid tumors. Results from the current real-world study indicate that the ChemoID assay is a valuable and practical tool for optimizing treatment selection when first-line therapy fails, and when there are multiple treatments available. The ChemoID assay takes 2–3 weeks to be completed from the date of receiving a live biopsy, which corresponds to the average time patients spend recovering from surgery before continuing further therapy. Therefore, the ChemoID assay is suitable for timely, individualized chemotherapy for cancer patients who received surgery. Furthermore, our results suggest that this individualized functional chemotherapeutic assay may indeed surpass the results achieved by empiric population-based treatment by providing better treatment options with improved outcomes. This compelling data suggests that the ChemoID CSC assay is beneficial in personalizing treatment strategies to increase survival time for recurrent GBM patients and to provide quality metrics for healthcare payers and providers to support access to care.


Patients with completely absent β-globin have a more severe clinical course, typically presenting in the first 6-24 months of life with severe anemia, failure to thrive, and end-organ damage.Routine transfusions result in excess iron accumulation, which can cause significant health consequences of their own, including pulmonary hypertension, liver dysfunction, and cardiac manifestations. The current standard of care for these individuals includes iron chelation therapy to help prevent these complications from occurring.5Historically, hematopoietic stem cell transplantation (HSCT) has been the only curative treatment available and is typically performed only in children younger than 14 years who have a human leukocyte antigen-matched sibling donor. Less than 25% of TDT individuals have access to compatible donors, which is a limitation of using HSCT as a curative treatment.Betibeglogne autotemcel  is a potentially curative gene therapy for β-thalassemia that utilizes a lentiviral vector to insert a functioning version of the HBB gene into the patient’s own red blood cells. Patients must also receive myeloablative chemotherapy prior to receiving beti-cel. The US Food and Drug Administration (FDA) accepted bluebird bio’s Biologics Licensing Application of beti-cel for priority review on November 22, 2021. The FDA Advisory Committee meeting was held in early June 2022, and the date by which a regulatory decision was due was set for August 19, 2022.https://icer.org/wp-content/uploads/2021/11/ICER_Beta-Thalassemia_Final-Report_071922.pdf.The Institute for Clinical and Economic Review (ICER) conducted a systematic literature review and cost-effectiveness analysis to evaluate the health and economic outcomes of beti-cel. Here we present the summary of our findings and highlight the policy discussion with key stakeholders held at a public meeting of the New England Comparative Effectiveness Advisory Council (New England CEPAC) on June 17, 2022. The detailed report is available on the ICER website: 10 Because of a manufacturing process modification following the phase 1 and 2 trials, which enhanced transduction efficiency in the phase 3 trials, evidence to inform our assessment of beti-cel was primarily derived from the 2 phase 3 trials (NorthStar-2 and -3), which enrolled 43 patients with TDT. Both NorthStar-2 and NorthStar-3 enrolled patients who were up to 50 years of age and received at least 8 transfusions per year for the last 2 years, or at least 100 mL/kg/year of packed red blood cell transfusions, and had genotypes consistent with TDT.7The evidence informing this review was derived from 2 phase 1 and 2 trials, 2 phase 3 trials, and 1 long-term follow-up study of trial participants.12 Over a median follow-up of 42 months (ranges 23-88) across phase 1 and 2 and phase 3 studies, no patients who achieved TI have lost TI.13The prespecified primary endpoint for the trials was transfusion independence (TI), defined as weighted average hemoglobin of 9 g/dL or more without any packed red blood cell transfusions for a continuous period of at least 12 months and beginning within 12 to 24 months of beti-cel infusion. Among 41 participants infused with beti-cel from the phase 3 trials, 37 (90%) achieved TI.14 Investigators deemed 2 cases of thrombocytopenia, 3 cases of abdominal pain, and 1 case each of leukopenia, neutropenia, mild thrombocytopenia, tachycardia, and pain in an extremity to be possibly related to the beti-cel infusion. There have been no deaths in patients with TDT across the beti-cel clinical development program, although serious AEs have been reported. All serious AEs were deemed likely related to myeloablative conditioning and included 5 cases of venoocclusive liver disease, stomatitis, thrombocytopenia, neutropenia, febrile neutropenia, pyrexia, and heart failure.15In the phase 3 trials, all patients who received beti-cel and its preceding myeloablative conditioning experienced at least 1 adverse event (AE).18Although there has been no evidence of insertional oncogenesis and no malignancies in the TDT trials of beti-cel, there have been reported cases of myelodysplastic syndrome and acute myeloid leukemia reported in gene therapy trials that have also used a lentiviral vector to treat other conditions.The small sample sizes of the trials create uncertainty around the estimates of some patient-important outcomes, particularly AEs. Some serious harms are likely rare occurrences and may not be observed in small trials. Other AEs such as infertility may require more than a decade to assess. In addition to a small sample, the length of follow-up (maximum 7 years) is reassuring but leaves residual uncertainty of the longer-term durability of treatment effect. And finally, the trials lacked a control group for reasons of ethics and feasibility; therefore, there is no head-to-head comparison of beti-cel with HSCT among patients who would be eligible for both.We developed a decision analytic model to evaluate the cost-effectiveness of beti-cel vs standard of care (transfusions and chelation therapy) in patients with TDT who are eligible for the gene therapy. Our analysis reported results from a health care system perspective and a modified societal perspective given that patient and caregiver productivity costs are high relative to direct health care costs and that the impact of beti-cel on costs outside the health care sector is substantial. The model used a lifetime horizon, and costs and outcomes were discounted at 3% per year.The model consisted of an upfront decision tree followed by a Markov model consisting of health states, including TD, TI, and dead. The percentage of patients achieving TI from beti-cel was based on trial data (90.2%), and the remainder of patients were assumed to remain TD over their lifetime. Redosing with beti-cel was not modeled. Given that the long-term durability of beti-cel is unknown, we assumed that starting in year 7, and each year thereafter, a small percentage of patients (0.271%) reverted to the TD health state . Patients reverting to the TD health state remained in that state until death.20 Full details of ICER’s cost-effectiveness analysis and model are available on ICER’s website: https://icer.org/wp-content/uploads/2021/11/ICER_Beta-Thalassemia_Final-Report_071922.pdf.The outcomes of interest included the cost per quality-adjusted life-year (QALY) gained, cost per LY gained, cost per equal value of an LY gained (evLYG), and TD years. The base-case analysis was modeled using a payment plan that had been publicly described by the manufacturer as its goal: a full upfront payment with an outcomes-based agreement consisting of an 80% payback option for patients who do not achieve TI. Based on additional public comments by the manufacturer, we assumed that the upfront price would be $2.1 million. In accordance with the modifications to the ICER value framework for ultrarare diseases and single and short-term therapies, we conducted scenario analyses, which included a 50/50 shared savings scenario analysis in which 50% of lifetime cost offsets from beti-cel are assigned to the health care system instead of being assigned entirely to the treatment.Results of the base-case analysis and the 50/50 shared savings scenario analysis are shown in Given the heterogeneity in patient preferences around an intervention like beti-cel, we emphasize that the model-projected lifetime health gains are best interpreted only within the subpopulation of those who would first consider that the patient-related benefits of beti-cel outweigh the harms. Estimating beti-cel’s lifetime health outcomes and costs was conducted under conditions of evidence uncertainty, including the risk of death from treatment and long-term durability of treatment effect. The modeled outcomes-based risk agreement, if administered, would reduce some of the impacts of this evidence uncertainty. The cost-effectiveness findings were most sensitive to the frequency of transfusions and the cost of chelator therapy, suggesting that if the costs of standard of care were largely reduced, then beti-cel’s value would also deteriorate.The New England CEPAC is one of the independent appraisal committees convened by ICER to engage in the public deliberation of the evidence on clinical and cost-effectiveness of health care interventions. The New England CEPAC comprises medical evidence experts, including practicing clinicians, methodologists, and leaders in patient engagement and advocacy. Their deliberation includes input from clinical experts and patient representatives specific to the condition under review, as well as formal comments from manufacturers and the public. A policy roundtable concludes each meeting, during which representatives from insurers and manufacturers join clinical experts and patient representatives to discuss how best to apply the findings of the evidence to clinical practice, insurance coverage, and pricing negotiations.The ICER report on beti-cel for β thalassemia was the subject of a New England CEPAC meeting on June 17, 2022. Following the discussion, the CEPAC members deliberated on key questions raised by ICER’s report. The panel voted 12 to 0 that the evidence was adequate to demonstrate that the net health benefit of beti-cel is superior to that provided by standard clinical management.The CEPAC panel also voted on “other potential benefits” and “contextual considerations” as part of a process intended to signal to policymakers whether there are important considerations when making judgments about long-term value for money not adequately captured in analyses of clinical and/or cost-effectiveness. They highlight several factors beyond the results of cost-effectiveness modeling that the CEPAC panel felt were particularly important for judgments of overall long-term value for money. The panel voted unanimously that beti-cel would have a major positive effect of offering a new treatment choice to patients, and the majority of the panel voted that beti-cel would have a major positive effect on patients’ and caregivers’ quality of life and their ability to achieve major life goals.As described in ICER’s Value Assessment Framework, questions on long-term value for money are subject to a value vote when incremental cost-effectiveness ratios for the interventions of interest are between $50,000 and $175,000 per QALY in the primary base-case analysis. All panel members voted that beti-cel was either intermediate (n = 3) or high (n = 9) long-term value for money when compared with standard clinical management. Full voting items and results are shown in https://icer.org/wp-content/uploads/2021/11/ICER_Beta-Thalassemia_Final-Report_071922.pdf. Several key policy recommendations for beti-cel are as follows: (1) All stakeholders have a responsibility to facilitate meaningful patient access to curative therapies for β-thalassemia in ways that do not exacerbate disparities. (2) Should the pricing and payment structure for beti-cel match the assumptions in this report, the treatment will meet traditional cost-effectiveness standards, and payers should therefore use the FDA label as the guide to coverage policy without seeking to unduly narrow coverage using clinical trial eligibility criteria. However, policy roundtable experts felt that it would not be unreasonable for payers to seek confirmation that patients do not have accessibility to a sibling-matched HSCT as first-line therapy. (3) Manufacturers should align prices with independent estimates of the patient-centered therapeutic value of their treatments. In the context of high-impact single or short-term therapies, transparent consideration should be given to a pricing scenario that “shares” any substantial cost-offset of treatment so that potentially large cost offsets are not used to justify exceedingly high 1-time prices.The policy roundtable discussion explored how best to translate the evidence and additional considerations into clinical practice and pricing and insurance coverage policies. The full policy recommendations can be found in the Final Evidence Report on the ICER website:


R2 = 0.169 to 0.266 versus 0.125 to 0.196; change in pseudo R2 = 0.055 to 0.087 versus 0.011 to 0.017). The temperament factors also explained more variance above and beyond the maladaptive factors when predicting the somatization factor . Although the HiTOP spectra measured by PID-5 are significant structures related to the categorical and dimensional measurements of somatoform, our findings highlight potential problems with both the construct and criterion validity of the HiTOP spectra.The Hierarchical Taxonomy of Psychopathology (HiTOP) is a phenotypic data-driven framework for the classification of psychopathology. We tested the construct and criterion validity of the HiTOP spectra measured by the Personality Inventory for DSM-5 (PID-5) using exploratory structural equation modeling (ESEM) and hierarchical regressions both to predict somatic symptom and related disorders (SSRD) and a somatization factor. The case–control study used hierarchical logistic regressions to distinguish 257 cases with SSRD from 1007 healthy controls by both the maladaptive and the temperament factors. The extracted factors were also used in hierarchical linear regressions to predict the dimensional somatization factor. The seven temperament factors explained more variance above and beyond the five maladaptive factors when predicting SSRD (pseudo  This structural data-based approach to the classification of general psychopathology uses factor analytical methods to cluster the mental symptoms into dimensional categories4. Compared to traditional diagnostic systems, HiTOP is a phenotypic model more consistent with both the genetic architecture of mental disorders and the impressions of environmental risk factors5 such as childhood abuse 7. This emerging classification system is also able to explain the long-term chronicity of psychopathology, account for functional impairment, and explain why disparate diagnoses from different classes respond to the same treatment patterns5.The Hierarchical Taxonomy of Psychopathology (HiTOP) is a data-driven structural framework of mental symptoms created to enhance the reliability and validity of clinical utilization8. This is a problem because of issues related to multifinality and equifinality. This means that mental symptoms do not always reflect etiology. It is possible for the same problem to be expressed by different symptoms, and for different problems to be expressed by overlapping symptoms. Thus, relying on symptom correlations may prevent a better understanding of the etiology of psychopathology, common etiological pathways responsible for the onset and progression of mental illness, and the dynamic developmental nature of psychopathological diagnoses9. At the conceptual level, HiTOP also implies “artefactual” comorbidity (e.g. patients with multiple disorders suffer from only one underlying condition) and a person-driven “snapshot” picture , and it neglects brain functional “perturbations” at the level of cell and systems. The instability of the higher-order structure over time, new inconsistencies caused by the extensions, and lack of explaining the rationale for the reintroduction of the somatoform diagnoses are considered at the methodological level9.Although there are some advantages over traditional diagnostic categories , HiTOP remains a phenotypic model that organizes psychopathology based on symptom correlations 1. However, recent research now provides additional empirical support for the somatoform as a distinct spectrum from the internalizing domain11. Despite the importance of the evidence, somatoform syndromes are not only differentially related to the gender and age group but are also associated with a wide range of spread and heterogeneous conditions in the HiTOP9. These issues and other problems with etiological pathways and genetic discovery are still considered serious challenges that complicate the implementation of the model, at least for somatoform diagnoses9.At first, HiTOP did not include the somatoform as a distinct category on the spectra level because there was not enough scientific evidence to support it12. Causal psychobiological mechanisms involved in somatoform syndromes include autonomic physiological arousal, the endocrine and immune systems, monoamine acids and neurotransmitters, brain mechanisms, and temperamental personality traits17. The association between some temperamental personality theories proposed by Akiskal et al.18, Cloninger20, and Lara et al.21, and somatoform diagnoses or Somatic Symptom and Related Disorders (SSRD) raised by the fifth edition of the Diagnostic and Statistical Manual of Mental Disorders (DSM-5) has already been investigated in some studies17. The five depressive, cyclothymic, hyperthymic, irritable, and anxious temperaments raised by Akiskal et al.18, are the affective components that were originally formulated for affective disorders. One study showed that cyclothymic, hyperthymic, irritable, and anxious temperaments are significantly related to somatization16. Cloninger's model suggests four temperament traits include novelty-seeking, harm avoidance, reward-dependence, and persistence20. Recent meta-analytic reviews have addressed strong associations between some temperaments such as harm avoidance and somatization15. Six emotional temperaments, including volition, anger, inhibition, sensitivity, control, and coping, along with twelve affective temperaments were components of Lara's dynamic theory21. A previous study reported associations between affective temperaments and somatic symptom severity in a large sample17.In contrast to the HiTOP symptom-focused model, various etiological models for somatoform syndromes highlight the interplay between cognitive-perceptual processes and behavioral, affective, and biological variants18, Cloninger20, and Lara et al.21. Because somatization is a complex psychiatric condition that is sometime introduced as an independent personality trait and that the dimensional personality assessment in patients with SSRD is widely ignored22, we first tried the extraction of a somatization factor  measured by various self-report instruments. We also tried to identify the higher-order factors of temperament traits measured by the Temperament Evaluation of Memphis, Pisa, Paris, and San Diego-Autoquestionnaire (TEMPS-A), the Temperament and Character Inventory (TCI), and the Affective and Emotional Composite Temperament Scale (AFECTS). Although there is little knowledge about the conjoint structure of temperaments and development in an integrated manner, personality traits are integrated within three disjoint genetic-environmental networks23.Given that temperamental personality theories, unlike the HiTOP, derive from etiological theories that rely on neurobiological structures involved in the development, we aimed to investigate both the construct validity of the HiTOP spectra measured using the Personality Inventory for DSM-5 (PID-5) and the criterion validity of the somatoform domain in a large non-Western sample by a comparison between the HiTOP spectra measured using the PID-5 and the temperamental predispositions proposed by Akiskal et al.24, is one of the instruments listed for measuring HiTOP constructs1. However, a dimensional personality disorder is strongly linked to the HiTOP structure. For example, a recent report addressed the role of the maladaptive personality domains assessed by PID-5 across HiTOP levels25. Another report attempts to address the placement of dimensional personality dysfunction within the HiTOP framework26. Although some studies in non-Western samples such as Iran have tried to test the construct validity of PID-5 maladaptive domains30, none of the studies included a large non-clinical sample as well as cases with SSRD15. Despite the present sample being from the west of Iran and this can make it difficult to generalize the results to other non-Western contexts, it is a unique effort to achieve advanced objectives. Finally, we tested the criterion validity of HiTOP by comparison between the maladaptive and temperament factors using hierarchical regressions to predict the somatization factor  and differentiate cases with SSRD from healthy controls .The second step was taken to investigate the construct validity of the HiTOP spectra by identifying the higher-order factors of PID-5. The PID-5, which was originally designed to measure personality pathologyThis case–control study included 257 cases with SSRD  and 1007 HCs  from the west of Iran. The samples were selected from the Kermanshah and Sanandaj cities between April 2020 and August 2021. The population of these urban areas is approximately 1.5 million individuals, the majority of whom belong to the Kurdish ethnic group. All samples were individuals aged 18 and above, unmedicated for mental illnesses within the past month, and fluent in the Farsi language. Samples of the control group were selected from the general population by public announcements on popular platforms using convenient sampling. The initial general population encompassed 1900 college attendees, employees at health science facilities and other educational establishments, individuals seeking medical assistance, and housewives. Initially, 82.8% of people completed and returned the questionnaires (n = 1581). Questionnaires of 214 people also contained 15 to 90% of the missing data (n = 1367). One hundred and one subjects were excluded from the study due to physical symptoms of coronavirus (n = 57), physical problems from other medical conditions (n = 29), multiple sclerosis (n = 8), hepatitis (n = 4), severe epilepsy (n = 1), cancer (n = 1), and drug addiction (n = 1). Finally, the data of 1266 people were found to be usable.32. Subsequently, a qualified clinical psychologist conducted an online diagnostic interview to examine the identified cases by the DSM-5 Criteria. Although 30 individuals were excluded due to their refusal to participate in the interview, the diagnosis of SSRD was verified for 229 individuals. As a result, the control group diminished from 1266 to 1007. Subsequently, we discovered 28 patients with SSRD from two psychiatric hospitals in Sanandaj and Kermanshah cities and appended them to the sample examined in the preceding phase (n = 28). Consequently, the number of cases with SSRD amounted to 257 individuals. All data and clinical interviews were carried out by two proficient psychologists. All subjects informed consent to participate in the study.In the following phase, individuals with SSRD were evaluated based on the self-reports of the subjects and specific threshold values (sensitive cutoff scores) for the Iranian population, which included scores exceeding 15.5 or 18 on the Screening for Somatic Symptom Disorders (SOMS-7) and the Short Health Anxiety Inventory (SHAI)24. Twenty five lower-order facets include emotional liability, separation insecurity, anxiousness, anhedonia, intimacy avoidance, withdrawal, grandiosity, manipulativeness, deceitfulness, irresponsibility, impulsivity, distractibility, perceptual dysregulation, unusual beliefs and experiences, eccentricity, risk-taking, callousness, attention-seeking, hostility, rigid perfectionism, perseveration, depressivity, submissiveness, restricted affectivity, and suspiciousness. The higher-order factors include negative affectivity, detachment, antagonism, disinhibition, and psychoticism. The domains are respectively equal to internalizing, detachment, antagonistic externalizing, disinhibited externalizing, and thought disorder on the HiTOP spectrum. Except for 16 items that are scored indirectly, each item is given a score between zero and three, from often false to often true. The higher scores on each domain or facet show a more severe pathology24. The PID-5 is a recommended instrument for measuring the HiTOP structure1, and its reliability and validity are acceptable in several Iranian samples30.The Personality Inventory for DSM-5 (PID-5) was used to evaluate the spectra on HiTOP. The PID-5 is a self-report dimensional questionnaire with 220 items to measure the pathological personality facets and five higher-order domains according to DSM-5 Section III33. The questionnaire measured five affective temperaments including depressive, cyclothymic, hyperthymic, irritable, and anxious traits. The Persian short version of TEMPS-A is formatted from 35 items for assessing depressive (8 items), cyclothymic (7 items), hyperthymic (8 items), irritable (6 items), and anxious (6 items) temperaments. All items of the questionnaire are scored directly , and the higher scores on each temperament trait (except hyperthymic) show more severe predispositions toward psychopathology. The Persian version of TEMPS-A is a reliable and validated instrument in the Iranian samples34.Temperaments were measured using three different instruments including the Temperament Evaluation of Memphis, Pisa, Paris, and San Diego-Autoquestionnaire (TEMPS-A), the Temperament and Character Inventory (TCI), and the Affective and Emotional Composite Temperament Scale (AFECTS). The original and short versions of TEMPS-A include 110 and 39 items, respectively35. Long forms of TCI are affected by cultural differences for Iranian samples36, and there is no official attempt to provide Iranian validations. Thus, we used the 125-item true–false version, which was well-standardized for Iranian culture with acceptable reliability and validity37. TCI-125 traits are novelty-seeking (20 items), harm avoidance (20 items), reward-dependence (15 items), persistence (5 items), self-directedness (25 items), cooperativeness (25 items), and self-transcendence (15 items). Sixty-one items in the questionnaire are scored directly  and 64 items are scored indirectly. Although the higher scores on the character components present more adaptive traits, very high/low scores on the temperament subscales show more severe predispositions toward psychopathology35. According to the objectives of the present study, we used only the temperament part.The TCI is a dimensional self-report tool with versions of 56, 125, 140, and 240 items designed to measure four temperaments and three character traits21. The Persian version of AFECTS is a reliable and validated instrument for the Iranian people38.The AFECTS is a two-part self-report scale with 60 items for dimensional measurements of the six emotional and twelve affective temperaments. The emotional section is a seven-point bipolar scale (score 0 to 7) for assessing volition (items 1–8), anger (items 9–16), inhibition (items 17–24), sensitivity (items 25–32), coping (items 33–40), and control (items 41–48). The emotional section is the personality dynamic part that tries to balance maladaptive predispositions  and adaptive predispositions . The affective section comprises internalizing , instable , stable , and externalizing  temperaments. Each item of the affective section is scored directly on a five-point Likert spectrum  and the higher scores show a more severe predisposition toward psychopathology39. The SOMS-7 has acceptable validity and reliability among Iranian normal and clinical samples31.Somatization and illness anxiety were assessed by the Screening for Somatic Symptom Disorders-7 (SOMS-7), the Patient Health Questionnaire-15 (PHQ-15), the Revised Form of Symptom Checklist-90 (SCL-90-R), and the Short Health Anxiety Inventory (SHAI). The SOMS-7 is a self-report questionnaire with 47 items for evaluating the severity of the somatic signs/symptom during the recent week. The questionnaire consists of four subscales including pain, cardiovascular and respiratory symptoms (17 items), gastrointestinal and urologic symptoms (17 items), neurological functioning symptoms (10 items), and musculoskeletal symptoms (3 items). Each item is scored directly on a four-point Likert scale  and the higher scores on the total questionnaire or subscales present more severe somatic symptoms40.The PHQ-15 is a self-report tool for measuring the severity of somatic symptoms during the last week. This questionnaire has no subscales and higher scores indicate more severe physical symptoms. Each of the items on the questionnaire is scored on a three-point scale from "not at all" (zero) to "a lot" (two) and the total score is ranged between zero and thirty. The reliability and validity of PHQ-15 have already been confirmed in Persian samples41. The Persian version of SCL-90-R is a reliable and validated instrument for the Iranian people42. In the present study, we used only the somatization subscale (12 items).The revised version of SCL-90 is a self-report questionnaire with 90 items for measuring mental symptoms. The nine subscales of the checklist include somatization, obsessive–compulsive disorder, depression, anxiety, hostility, phobic anxiety, interpersonal sensitivity, paranoid ideation, psychoticism, and six additional items. Each of the items scored on a five-point Likert scale from "no discomfort" (zero) to "very severe discomfort" (four) and the higher scores on the subscale or total questionnaire present a more severe mental symptomatology32. The reliability and validity of SHAI have already been confirmed in Persian samples43.The SHAI is a self-report questionnaire with 18 items for assessing health anxiety. The three subscales of the questionnaire include rumination (7 items), probability of disease (7 items), and negative outcome (4 items). Each item of the SHAI is scored directly on a four-point Likert scale from "very low" (zero) to "very high" (three). The total score on the questionnaire is between zero and 54 and the higher scores show more severe health anxiety45. We compared model fit for all factor solutions using Aikake Information Criteria (AIC), Bayesian Information Criteria (BIC), Root-Mean-Square Error of Approximation (RMSEA), Comparative Fit Index (CFI), and Tucker-Lewis Index (TLI) for four- to sex-factor solutions of the PID-5 as well as three- to eight-factor solutions of the temperament measures. A CFI equal to or greater than 0.96 and RMSEA equal to or less than 0.06 were acceptable for the select number of the factors47. We chose the parsimonious model including less latent factors when a change in CFI was less than or equal to 0.0148. Maximum likelihood estimations with Oblimin rotations performed by Mplus 7.449 were used to select the traits/temperaments with the strongest loading on the factor-of-interest and the weakest loadings on all other factors.In the first stage, we conducted a conjoint Exploratory Factor Analysis (EFA) with maximum likelihood estimations on four somatization measures including SOMS-7, PHQ-15, SCL-90-R somatization, and SHAI scores. Because we found a strong pattern of intercorrelations for both PID-5 and temperament measures, we test exploratory structural equation modeling (ESEM) to examine the factor structures of both the PID-5 and the temperament measures. ESEM is an effective approach that offers a confirmatory test of the early factor structures while allowing for the estimation of all cross-loadingsR2 to determine how much additional variance each model was explained in the outcome. The validity of the factors related to SSRD in each block was reported using Cox & Snell along with Nagelkerke (pseudo R2). The factors correlated to the somatization factor were identified using Beta statics and the factors related to SSRD were identified using Wald statics and an odd ratio (OR) with a 95% confidence interval (CI). All analyses were performed using SPSS software version 20 and a significance at the level of less than 0.05 was considered.In the next step, we used factor scores from ESEM analyses as predictors to construct two hierarchical linear regression models (somatization factor) and two hierarchical logistic regression models (SSRD group), in which the derived factors from both the PID-5 and temperament measures were alternatively entered as blocks to predict both the somatization factor and cases with SSRD, respectively. We compared the change in All subjects informed consent to participate in the study. This study is consistent with the Helsinki guidelines and it was approved by the ethics committee of the Kurdistan University of Medical Sciences (IR.MUK.REC.1398.169).χ2 = 18.576, p < 0.001). The SOMS (= 0.867), PHQ (= 0.860), SCL90 somatization (= 0.831), and SHAI (= 0.444) respectively loaded more strongly on the extracted factor.The mean, standard deviation, skewness, kurtosis, and Cronbach's alpha of the somatization measures in the full sample (n = 1264) are seen in Appendix Table . The conBoth PID-5 and temperament measure statistics including the mean, standard deviation, skewness, kurtosis, and Cronbach's alpha of the scale items in the full sample are seen in Appendix (Tables S2 & S3). The mean score of all the PID–5 subscales was ranging from 0.70  to 1.30 (risk-taking) while the mean score of all the temperament subscales was ranging from 1.40 (TEMPS-A anxious) to 37.03 (AFECTS coping). Cronbach’s alpha of all the PID–5 subscales were ranging from 0.51 (suspiciousness) to 0.91 (depressivity and eccentricity), while Cronbach’s alpha of all the temperament subscales were ranging from 0.43 (TCI reward dependence) to 0.91 (AFECTS volition). p < 0.001, change in CFI > 0.01). Table We compared three ESEM models four- to eight-factor solutions to test the structure of the temperament measures, the results of which are shown in Table  p < 0.001, change in CFI > 0.01). Table We compared three ESEM models four- to sex-factor solutions to test the structure of the PID-5, the results of which are shown in Table R2 of the model was 0.269 (p < 0.001), whereas when the seven temperament factors were entered in the first block, the R2 of the model was 0.392 (p < 0.001). As such, our main focus is on the relative change in R2 values for models with the somatization factor as the dependent variable. The temperament factors were explained more variance above and beyond the maladaptive factors when predicting the somatization factor . Conversely, the maladaptive factors were explained less variance above and beyond the temperament factors . Overall, all the factors could predict a 41.5% variance of the somatization factor). We also report standardized Beta coefficients from each of the specific factors, which the results also can be seen in Table β =  − 0.270 and 0.335, p < 0.001) and two temperament factors  are significantly related to the somatization factor. Therefore, the factors of both systems are complementary to predict the somatization factor.Table R2 of the model was 0.125 to 0.196 (p < 0.001), whereas when the seven temperament factors were entered in the first block, the pseudo R2 of the model was 0.169 to 0.266 (p < 0.001). As such, our main focus is on the relative change in R2 values for models with SSRD as the dependent variable. The temperament factors were explained more variance above and beyond the maladaptive factors when predicting SSRD . Conversely, the maladaptive factors were explained less variance above and beyond the temperament factors . Overall, all the factors predict an 18 to 28.3% variance of SSRD. We also report Wald statistics and odd ratio (OR) with a 95% confidence interval (CI) for each of the specific factors, which the results also can be seen in Table , p = 0.009) and the two temperament factors  are significantly related to SSRD. Therefore, the factors of both systems are complementary to differentiate cases with SSRD from healthy controls.Table 10. Although more recent studies attempt to restore the validity of HiTOP and useful tools using the SSRD raised by the DSM-550, convincing cross-cultural empirical evidence with a prospective longitudinal approach in large populations, especially non-Western samples, is not yet available. These considerations highlight the necessity of validating the HiTOP model as currently, there is insufficient evidence to introduce the somatoform as an independent spectrum on HiTOP9.We examined the validity of HiTOP by comparison between the maladaptive and temperament factors, both to predict SSRD and the somatization factor. Unlike temperamental theories, the HiTOP is a phenotypic  classification system that is agnostic to neurobiological pathways and other etiological causal mechanisms related to psychopathology. The somatoform spectrum on HiTOP also relies mainly on findings based on the DSM-IV diagnostic criteria or earlier versions of DSM for somatoform diagnoses and similar categories such as hypochondriasis51. The difficulty for generalizability is the most important limitation of the developed models, which can be well tested using construct validation across cultures before clinical application52. We assessed both construct and criterion validity using factor analysis and regression techniques. Although factor analysis and ESEM techniques test construct validity by identifying strongly correlated latent factors in a dataset, the higher-order common factors only characterize the phenotypic aspects of psychopathology. Such a basis cannot provide accurate insight into the effects of genotypic and environmental factors on the incidence and development of psychiatric disorders such as somatization.The purpose of the current study was to further investigate both the construct and criterion validity of the HiTOP spectra. This is important not only for testing the validity of HiTOP but also because the validation of psychological models helps to improve the illness screening system and provide appropriate interventions to prevent more serious outcomes. The validation of models can be done internally using the same source or comparing explanations with other models in new contexts and externally using new data from an independent data set53. However, recent studies in Iran showed that disinhibition traits tend to load on the negative affectivity and antagonism factors29. In our study, some disinhibited traits including distractibility and impulsivity were strongly loaded on the negative affectivity factor, while irresponsibility and risk-taking were loaded on the antagonism factor. The narcissism factor containing anankastic (rigid perfectionism) and antagonistic (attention seeking and grandiosity) traits is somewhat similar to the results of the previous report in the samples of western Iran28. Therefore, we conclude that the replicability of the maladaptive domains of the PID-5 and the HiTOP spectra in the present non-Western sample is somewhat questionable. Despite supporting the five-factor structure by ESEM, not extracting the disinhibition factor independently can indicate the challenges of the construct validity of spectra on the HiTOP. However, the difference in factor loads is probably due to the estimation and rotation methods54.Our results showed that PID-5 facets loaded on five independent higher-order factors including the components of negative affectivity or internalizing (Factor I), antagonistic externalizing (Factor II), detachment (Factor III), thought disorder (Factor IV), and narcissism (Factor V). Although factors I to IV cover all domains of the HiTOP spectra, the fifth extracted factor do not correspond to the five independent factors introduced by both the PID-5 and HiTOP. Not the disinhibition factor but the narcissism factor including attention seeking, grandiosity, rigid perfectionism, and hostility was identified. This finding was not consistent with the results of recent meta-analyses that include the disinhibition factor21. Our findings showed that temperament traits, loaded on seven independent higher-order factors, including instability (Factor I), negative affectivity (Factor II), positive emotionality (Factor III), negative emotionality (Factor IV), internalization (Factor V), externalization (Factor VI), and stability (Factor VII). Previous findings support the hierarchical structure of temperaments proposed in different theories56. Although we could not find studies to identify the higher-order factors of the three proposed theories, a recent report showed that temperament and character traits are integrated within some unique genetic-environmental networks23.We also aimed to identify higher-order temperament factors of the 27 traits proposed by three independent theoriesThe findings of the present study show that the higher-order factors extracted from PID-5 and temperament predispositions can potentially complement each other in the categorical and dimensional measurement of the somatoform spectrum on HiTOP. This finding supports the discriminant validity of both maladaptive and temperament factors. However, the incremental validity of maladaptive factors was much smaller than the temperament factors. In more detail, the PID-5 maladaptive factors in both the categorical and dimensional assessments of somatization were able to improve the temperament factors by only two percent. Although the partial incremental contribution was statistically significant, it was much smaller than the incremental validity of the temperament factors . Although we know that adding more predictors to a regression model may provide more explanation for the variance, we aimed to compare maladaptive and temperament factors to identify a more efficient framework.1. The PID-5 facets are less involved in the underlying psychopathology of somatoform disorders, including attachment style, emotional awareness and dysregulation, and alexithymia61. Although emotional dysfunction was recently introduced as a higher-order factor covering somatoform and internalizing domains10, the PID-5 only includes a few underlying and revealing facets of the somatoform such as emotional liability, separation insecurity, anxiousness, perceptual dysregulation, restricted affectivity, and distractibility (low)63. Following such problems, recent efforts have been directed to design practical tools with the ability to measure various structures on all levels of HiTOP64. But unlike the recommended tools for measuring HiTOP structures, temperament theories and tools, especially emotional and affective temperaments, are completely dependent on emotional structures predisposing to psychopathology21. The results of the present study indicated that both the negative affectivity factor and internalization factor of temperament are facilitators of both somatization and SSRD. Some previous evidence also supports the key role of emotional and affective temperaments in SSRD and somatization65. Our conjoint factor analytic measures are useful for testing the cross-cultural generalizability of the extracted constructs. However, the present research do not have the merits for cross-cultural studies of etiology and development that are based on measures of individual differences in biological and genetic variables that are shared by all humans across cultures and environments, as validated by TCI temperaments67.These findings underscore potential problems with the criterion validity of the HiTOP spectra, at least about the intercorrelations with the somatoform spectrum. Of course, it should be noted that it is not practically possible to measure all spectra structures or other HiTOP levels even with recommended tools such as PID-568. Thus, the current study provides much-needed diversification to the research field. However, we only included samples from western Iran (mainly Kurdish culture), which makes it difficult to generalize the findings to other regions of Iran or other non-Western contexts. We tried to achieve advanced findings about a more comprehensive classification of temperaments by identifying higher-order temperament factors proposed in three independent theories21. Dimensional and categorical measurement of somatoform in the role of criterion variable provided the possibility of independent analyses, and similar findings related to both assessments supported the valid methodology and analysis of the present study. We reported the internal consistency of all questionnaires and found all of them to be reasonably reliable. This supports the validity of the assessment and reduced bias caused by the self-report instruments used in the present cross-sectional study. Conversely, using one tool such as PID-5 to evaluate the constructs of a model such as HiTOP is not enough, and validation of the psychological models requires the use of more comprehensive tools64. Using self-report questionnaires at one point in time only provides a person-driven “snapshot” picture9, while prospective longitudinal studies to investigate the external temporal validation51 can provide valuable information. Although comorbidity cannot confound the results related to the somatoform dimensional measurement, it may potentially affect the findings obtained from the categorical assessment of the case–control study. Finally, the gender distribution of the sample in the two groups was somewhat unequal. The lack of statistical significance of this difference should not mislead us and we recommend gender matching in future studies.Our work with a sectional case–control study and the large sample size is unique, at least in a non-Western context. Indeed, The target populations in most clinical psychology studies are white people from the West4. We found that the spectra on HiTOP measured by PID-5 are significant structures related to both the categorical and dimensional measurements of the somatization. Our findings also support the discriminant validity of both maladaptive and temperament higher-order factors as potentially complementary structures associated with the SSRD. Nevertheless, the incremental validity of maladaptive factors to both the categorical and dimensional somatization was much smaller than the temperament factors. These results highlight potential problems with the criterion validity of the HiTOP spectra, at least about the intercorrelations with the somatoform spectrum. Future studies considering the methodological limitations of the present study can provide more evidence for the validity of the spectra on HiTOP.In sum, evidence from the present study supports a five-factor structure relatively similar to the PID-5 containing a narcissism factor instead of a disinhibition factor. This indicates a relatively weak construct validity, which, in addition to culture, can be partially influenced by the estimation and rotation methods. These results also raise further questions about the generalizability of HiTOP to people from the non-Western world. The HiTOP framework is derived almost exclusively from data from American and European samples, and thus, may not generalize (or it may even promote systematic bias) to other cultures and underrepresented groupsSupplementary Tables.