This is a prognostic modeling study with a retrospective cohort design. We used the TRIPOD+AI (Transparent Reporting of a Multivariable Prediction Model for Individual Prognosis or Diagnosis + Artificial Intelligence) guidelines for reporting our work (Checklist 1) [31]. This study used data from OneFlorida+, a secure and centralized repository of patient‐level data from both health system partners and insurers, and managed by the OneFlorida Clinical Research Consortium [32]. Currently, OneFlorida+ includes EHRs and claims data from 26 million individuals across Florida, Georgia, Alabama, and Arkansas, covering a wide range of populations and health care settings [30,33]. All data sources are mapped to the Patient‐Centered Outcomes Research network common data model (version 7.0) to ensure standardization of data elements across sources. Major data elements in the common data model include demographics, enrollment, encounters, diagnoses, procedures, dispensed medications, and deaths. Additionally, while OneFlorida data includes Florida Medicaid recipients with claims data only, our study cohort requires having at least 1 EHR encounter. The study population included adult patients (aged ≥18 y) who had at least 1 opioid prescription filled between 2017 and 2022 (Figure S1 in Multimedia Appendix 1). The index date was defined as the date of a patient’s first opioid prescription between October 1, 2017, and September 30, 2022. To construct a cohort appropriate for assessing incident OUD risk, we excluded patients who (1) had cancer diagnoses (Table S1 in Multimedia Appendix 2), (2) received hospice care, (3) had a diagnosis of OUD, opioid overdose, other substance use disorders, drug abuse, or received methadone or buprenorphine for OUD before initiating opioids, or (4) did not have at least a 3-month observation window following the first opioid prescription to allow for predictor measurement. We also excluded patients with a diagnosis of other substance use disorders to minimize confounding, as some physicians may have used this diagnosis for patients with co-occurring OUD and other substance use conditions. Once eligible, patients remained in the cohort until they experienced an outcome of interest (ie, incident of OUD diagnosis) or were censored due to death, regardless of continued opioid use (Figure S2 in Multimedia Appendix 1).

For external validation, we applied the algorithm to 2018‐2020 EHR data from the UPMC health system in Pennsylvania, a region with demographic and opioid prescribing characteristics distinct from the OneFlorida+ network. Cohort construction and predictor generation followed the same procedures used for the OneFlorida+ dataset.

Similar to other claims-based analyses [29,34-36], our primary outcome was incident OUD (Table S2 in Multimedia Appendix 2), defined as the first recorded diagnosis of OUD from all settings in each 3-month window after the index prescription date (Figure S2 in Multimedia Appendix 1). The ICD-10 (International Statistical Classification of Diseases and Related Health Problems 10th Revision) codes for incident OUD diagnosis include F11.1X and F11.2X but exclude F11.11 (opioid-related disorders in remission) and F11.21 (opioid dependence in remission).

We identified 183 candidate predictors of OUD from the prior literature (Table S3 in Multimedia Appendix 2) [37-48]. These predictors included sociodemographic factors, patient health status, and patterns of opioid use and other nonopioid prescriptions measured at baseline (during the 3 month period before the first opioid fill) and in 3-month windows after initiating prescription opioids. This 3-month window was selected for both clinical and operational relevance. Evidence shows that early indicators of problematic opioid use (eg, dose escalation, early refills, or polypharmacy) typically emerge within the first 3 months of therapy [49-52]. Pharmacoepidemiologic studies demonstrate that transitions to sustained or high-risk opioid use most commonly occur within 3 months after treatment initiation [53,54]. This timeframe also aligns with the quarterly monitoring cycles used by many prescription drug monitoring programs and health plans [14,38]. We updated the predictors measured in each 3-month period to account for changes over time for predicting incident OUD risks in each subsequent period (Figure S2 in Multimedia Appendix 1). This time-updating approach mimics active surveillance that a health system might conduct in real time [55] to provide clinicians with timely, actionable opportunities (eg, closer follow-up, patient counseling, or medication adjustments) before the progression to OUD. We examined missingness across candidate predictors and applied prespecified imputation rules before model training. We used simple imputation techniques where missing values for all categorical predictors (including race, ethnicity, and provider sex) were imputed using the modal category. Continuous predictors (eg, use counts, comorbidity scores, and medication counts) with missing values were imputed using the median calculated from the full analytic dataset. Binary and count predictors representing the presence or frequency of diagnoses, procedures, or medications were imputed as zero when missing, consistent with the interpretation that the absence of a code reflects no recorded event during the baseline window.

Our machine learning analysis using OneFlorida+ data had two objectives: (1) developing a prediction model to generate individuals’ incident OUD risk scores and (2) stratifying individuals into subgroups based on similar OUD risk levels. We developed and tested 4 ML algorithms to predict incident OUD: elastic net (EN), least absolute shrinkage and selection operator (LASSO), gradient boosting machine (GBM), random forest (RF). These algorithms have consistently demonstrated strong predictive performance in prior studies [12,46,56,57]. Details of each algorithm are provided in Multimedia Appendix 3. We randomly divided the cohort into 3 subsets: a training sample for algorithm development, a testing sample for algorithm refinement, and a validation sample for evaluating predictive performance. This split was conducted strictly by patient ID, such that all 3-month episodes from a given patient were assigned to a single dataset (training, testing, or validation), and no patient contributed episodes to more than one set, thereby avoiding information leakage across datasets. Because patients could contribute multiple 3-month episodes, episode-level observations within patients were correlated by design. Models were trained and evaluated at the episode level to reflect the intended real-world use case of repeated, longitudinal risk assessment over time and to maximize learning from rare outcomes. Tree-based ensemble methods such as GBMs are generally robust to correlated observations because they do not rely on independence assumptions and learn predictive structure through recursive partitioning rather than parametric estimation. To evaluate the potential impact of within-patient correlation on model performance estimates, we conducted sensitivity analyses using patient-level random subsets. We used one 3-month period with predictor measurements to forecast risk in the subsequent 3 months within the validation set.

To evaluate discrimination performance, we assessed whether individuals predicted to be at high risk had higher OUD incidence than those predicted to be at low risk. We compared C-statistics across different methods in the validation sample using the DeLong test [58]. A C-statistic between 0.7 and 0.8 indicates good discrimination, while values above 0.8 indicate very good discrimination. Precision-recall curves were also examined [42]. Given that OUD events are rare outcomes and C-statistics do not incorporate information about outcome incidence, we reported other metrics, including sensitivity, specificity, positive predictive value (PPV), negative predictive value (NPV), number needed to evaluate (NNE) to identify 1 OUD episode, and estimated rate of alerts, to thoroughly assess the algorithms’ predictive ability (Figure S3 in Multimedia Appendix 1) [43,59].

There is no universally applicable prediction probability threshold, so we evaluated performance across multiple sensitivity and specificity levels (eg, selecting 90% sensitivity). To determine an optimized threshold that balances sensitivity and specificity, we applied the Youden index in the training sample [60]. In the validation sample, patients were stratified into risk subgroups based on deciles of predicted incident OUD probability. The highest decile was further divided into 3 strata: the top first percentile, the second to fifth percentiles, and the sixth to tenth percentiles. This approach allowed for a more detailed assessment of individuals at the highest risk of developing OUD.

We conducted additional analyses to enhance clinical use. The primary goal of our ML algorithm was to generate risk scores for incident OUD. For comprehensive model explanation and further enhancement of model interpretability, we conducted a Shapley additive explanations (SHAP) analysis [61]. SHAP quantifies the contribution of each predictor to the model’s output while accounting for interactions with other features. The SHAP plots indicate whether the mentioned features have a positive correlation (red color) or a negative correlation (blue color) with OUD. To evaluate bias across demographic factors (eg, race, age, and sex), we examined risk score distributions and false positive rates (FPR) and false negative rates (FNR) in the best performing model. We defined significant disparity as a parity ratio between subgroups falling outside the range of 0.80 to 1.25, consistent with the standards used in algorithmic fairness audits [62].

Additionally, we evaluated model calibration to assess the reliability of predicted risk scores. Given the potential for miscalibration in rare-outcome ML models, we applied isotonic regression to recalibrate predicted risk scores [63]. The overall accuracy of probabilistic predictions was assessed using the Brier score metric [64]. To determine clinical use beyond discrimination, we performed decision curve analysis. We calculated the net benefit [65] across a range of clinically relevant risk thresholds, comparing the recalibrated model against default strategies of “treat all” and “treat none”. This approach quantifies the trade-off between the benefit of identifying true OUD cases and the potential harm of false-positive alerts.

All analyses were performed using SAS 9.4 (SAS Institute Inc) and Python 3.6 (Python Software Foundation). Continuous variables were summarized using mean (SD) values, while categorical variables were summarized using frequencies and percentages. We compared patients’ characteristics in training, testing, and validation datasets using 2-tailed Student t tests, chi-square tests, and ANOVA, or corresponding nonparametric tests.

This study was reviewed and approved by the University of Florida (UF) Institutional Review Board (IRB202101897). The research involved secondary analysis of limited electronic health record data from the OneFlorida+ Data Trust. In accordance with federal regulations (45 CFR 46.104[d] [4]) and institutional policies, the requirement for informed consent was waived because the study involved no direct participant contact and posed minimal risk to individuals. All analyses were conducted within the secure UF server environment; no identifiable private information was accessed or transmitted outside the system. The study adhered to institutional data-use agreements and complied with all relevant guidelines for human subjects research and data privacy. The external validation using UPMC data was reviewed by the University of Pittsburgh Human Research Protection Office and determined not to constitute human subjects research, as only deidentified data were used. Therefore, institutional review board review and informed consent were not required.

OneFlorida+ patients in the training (n=60,694), testing (n=60,695), and validation (n=60,694) samples had similar characteristics and outcome distributions (mean age 53.02, SD 7.60 y; 111,576/182,083, 61.28% female; 125,208/182,083, 68.76% White; 144,316/182,083, 79.26% non-Hispanic; 4328/182,083, 2.38% had incident OUD; Table 1). The external validation cohort from the UPMC included 129,215 patients (mean age 58.91, SD 16.75 y; 78,574/129,215, 60.81% female; 115,907/129,215, 89.7% White; 123,083/129,215, 95.25% non-Hispanic; 2537/129,215, 1.96% had incident OUD). While sex distributions were similar between the 2 health systems, the UPMC cohort was older and less diverse, with more White and non-Hispanic patients.

Figure 1 summarizes the 4 prediction performance measures of each model in the internal validation sample. GBM (C-statistics=0.879, 95% CI 0.874‐0.884) and EN (C-statistics=0.872, 95% CI 0.867‐0.877) models outperformed the LASSO (C-statistics=0.846, 95% CI 0.840‐0.851) and RF (C-statistics=0.798, 95% CI 0.792‐0.804) models, with the GBM model requiring the fewest predictors (n=75) for predicting 3-month incident OUD (P<.001; Figure 1A). The GBM and EN models had similar prediction performance, and the GBM model had the best precision-recall performance (Figure 1B). Sensitivity analyses using patient-level data yielded similar results (Figure S4 in Multimedia Appendix 1). To evaluate model robustness across key clinical subgroups, we stratified the GBM model’s performance by age, sex, and ethnicity. The model demonstrated consistent discriminative ability across populations, with C-statistics ranging from 0.833 to 0.909 across age groups (Figure S5 in Multimedia Appendix 1), 0.865 to 0.887 between sexes (Figure S6 in Multimedia Appendix 1), and 0.864 to 0.914 by ethnicity (Figure S7 in Multimedia Appendix 1).

Table S4 in Multimedia Appendix 2 shows the performance measures for predicting incident OUD across different levels (90%‐100%) of sensitivity and specificity for GBM and EN. When set at the optimized sensitivity and specificity as measured by the Youden index, the GBM model had a sensitivity of 76.59%, specificity of 84.75%, PPV of 4.36%, NPV of 99.75%, and NNE of 23; and the EN model had a sensitivity of 77.65%, specificity of 82.74%, PPV of 3.93%, NPV of 99.76%, and NNE of 25 (Figures 1C and 1D; Table S4 in Multimedia Appendix 2).

Figure 2 represents the actual OUD rate for individuals in each decile subgroup using GBM. The high-risk subgroup (with risk scores in the top decile; 11% [n=6696] of the validation cohort) had a PPV of 3.26%, a NPV of 99.8%, and NNE of 31. In a hypothetical clinical application involving 1000 patients, applying the top decile threshold would generate alerts for approximately 100 individuals. Based on the observed PPV at this threshold, approximately 4 of the 100 flagged patients would be expected to develop OUD, demonstrating the model’s ability to meaningfully concentrate risk among a small subgroup. Conversely, the high NPV (99.8%) indicates that very few incident OUD cases would arise among the 900 unflagged patients, consistent with the model’s low false negative rate. Taken together, these estimates illustrate how the model could support prioritization of patients most likely to benefit from closer monitoring or preventive intervention.

Among all 323 individuals with an incident OUD, 265 (82.04%) occurred in the top two decile subgroups (Decile 1=67.5 % and Decile 2=14.6%). Those in the first decile subgroup had at least a 14-fold higher OUD rate compared to the lower-risk groups (eg, observed OUD rate: Decile 1=9.51%, Decile 2=0.67%, Decile 10=0.02%). The third through tenth decile subgroups had minimal OUD incidence rate (2 to 16 per 10,000).

In our evaluation of racial fairness, we found that among the true positive cases identified in the top decile risk subgroup by the GBM model, 78.44% (171/218) were White individuals and 18.8% (41/218) were Black individuals. In the top first percentile, the distribution was 67.3% (35/52) White individuals and 25% (13/52) Black individuals. The detailed racial distribution of true positives across all risk subgroups is provided in Table S5 in Multimedia Appendix 2. We also compared FNR and FPR by race in the GBM model with and without race as a predictor. FNRs were consistently higher for Black individuals compared to White individuals across most percentiles, while FPR remained similar across racial groups (Figure S8 and Figure S9 in Multimedia Appendix 1). We compared the FNR and FPR between racial groups at the 90th percentile and found that the FNR parity ratio (1.23) and FPR parity ratio (1.03) were within the accepted threshold (0.8‐1.25) [62], suggesting equitable model performance across racial groups when race was included as a predictor. We also assessed fairness across age and sex. The true positive cases in the top decile were concentrated among the 35‐ to 50-year age group (82/218, 37.61%) and females (146/218, 66.97%). The detailed distribution of true positives by age and sex across all risk subgroups is provided in Tables S6 and S7 in Multimedia Appendix 2. FNRs were slightly higher for older individuals (≥65 y) across most percentiles compared to the young adults (18‐34 y), while FNRs for males and females were closely aligned. FPRs remained consistent across both age groups and sex (Figures S10 and S11 in Multimedia Appendix 1). Parity ratios for sex (FNR=0.91, FPR=1.18) and age-based FPR (0.91; comparing 18‐34 vs ≥65 y) remained within the acceptable range (0.8‐1.25), whereas the age-based FNR parity ratio (0.74) was below the threshold.

Figure 3 presents variable importance plots using SHAP values derived from the GBM model. The plots on the right-hand side present the average impact of various features on incident OUD, listed in descending order of their significance. Variables such as age and history of other pain for each 3-month period were negatively correlated with incident OUD, while factors having a positive correlation with incident OUD included the number of outpatient visits, history of back pain, and Elixhauser comorbidity index score.

Calibration assessment showed that the GBM model initially overestimated risk (Brier score=0.09), a known phenomenon in rare-event prediction. After isotonic recalibration, model reliability improved significantly, with a calibration slope of 0.98 (ideal=1), an intercept of –0.03 (ideal=0), and a Brier score of 0.01 (lower is better; Figure S12 in Multimedia Appendix 1). Decision-curve analysis showed that a “treat-all” strategy (intervene on everyone) yields negative net benefit at risk thresholds above 0.5% (Figure S13 in Multimedia Appendix 1), whereas the GBM model maintained positive net benefit across a broader range of clinically relevant thresholds. For example, at the top decile cutoff (probability threshold=0.009), the model achieved a net benefit of 0.0026. This corresponds to identifying approximately 2.6 additional true-positive OUD cases per 1000 patients (after accounting for false-positive alerts) compared with a strategy of no intervention. Relative to the OUD outcome incidence (0.53% in the prediction window), this net benefit reflects ~50% of the maximum achievable use for a perfect classifier in this setting, indicating substantial clinical value despite the low-prevalence context.

When applied to the UPMC cohort, the best-performing GBM model achieved a C-statistic of 0.756 (95% CI 0.750‐0.762; Figure S14 in Multimedia Appendix 1). The high-risk decile subgroup demonstrated a PPV of 2.52%, an NPV of 99.8%, and NNE of 40 (Figure S15 in Multimedia Appendix 1). The top first percentile had an OUD incidence rate 14-fold higher than the overall baseline (8.15% vs 0.58%). Fairness analyses in the UPMC cohort revealed variations in FNR and FPR across age groups (Figure S16 in Multimedia Appendix 1), while error rates remained similar across race (Figure S17 in Multimedia Appendix 1) and between males and females (Figure S18 in Multimedia Appendix 1). For example, using the 90th percentile risk score as the elevated risk cutoff, younger adults (18‐34 y) had a lower FNR (fewer missed cases) compared to older adults (≥65 y).

This study expanded our previous work using ML approaches to improve the accuracy of predicting incident OUD in the subsequent 3 months of prescription opioid initiation among fee-for-service Medicare beneficiaries and broaden the applicability of these models across a diverse population in the OneFlorida+ network [29]. Our GBM and EN models demonstrated strong predictive performance, achieving a high C-statistic (>0.87) and outperforming LASSO and RF. Our best-performing GBM model offers several advantages, including its ability to eliminate the need for a separate feature selection process and its flexibility in hyperparameter tuning to capture complex interactions between predictors and outcomes. We acknowledge that this flexibility in model tuning can be computationally intensive and time-consuming. The algorithm effectively stratified the population into distinct risk groups based on predicted risk scores. Notably, approximately 80% of the cohort had minimal OUD risk, while the highest-risk decile alone accounted for ~68% (218/323) of all individuals who developed OUD. Given the low incidence of OUD within a 3-month period, the PPV was expectedly low. However, in the context of OUD prevention, a lower PPV may be ethically and clinically acceptable when associated interventions are supportive (eg, closer follow-up, motivational interviewing, review of opioid therapy, or other high-risk concurrent medication use) rather than punitive actions (eg, refusal to prescribe or abrupt discontinuation). For these low-risk, supportive interventions, the potential harm of a false positive (providing additional support to a patient who would not have developed OUD) is minimal compared with the harm of missing a true high-risk case (false negative). It is therefore important to frame the model as augmenting, not replacing, clinician judgment to ensure that low PPV does not lead to overreaction to false positives or unintended consequences.

A key strength of this study is the successful external validation of our algorithm using an independent UPMC cohort. Our model maintained good performance (C-statistic: 0.76) without retraining, demonstrating transportability across two distinct health care systems. This replication supports the stability of the key predictors across settings and indicates that the model can effectively identify high-risk individuals in diverse clinical environments.

Prior studies, targeting different aspects of OUD risk, varied in their prediction windows and data sources, including 6-month OUD risk based on private insurance claims [66]; 12-month risk of aberrant opioid use behaviors following an initial pain clinic visit [67]; and 12-month OUD risk using private insurance claims [38,68] or pharmacy benefit manager claims data [36]. Other models focused on longer-term risk, such as 2-year problematic opioid use documented in EHR from primary care [69] and 5-year OUD risk using electronic medical record from a medical center [34] or Rhode Island Medicaid data [35]. Despite their contributions, these models had key limitations. Most measured predictors are only at baseline, lacking temporal updates over time. Several relied on case-control designs, which may not generalize well to population-level OUD incidence rates. Additionally, even in non–case-control designs, the highest reported C-statistic was 0.85, indicating room for improvement in predictive performance [35,36,67,69]. Expanding on our prior Medicare claims-based study, this work explored whether predictive performance can be refined using EHR data. Our study addressed the abovementioned limitations by using a population-based cohort and a more immediate prediction window, estimating OUD risk within the next 3 months rather than over a year or longer. This short-term predictive approach, combined with innovative risk stratification and enhanced clinical interpretability, made our method highly applicable for timely intervention.

Although the findings of the study are promising, some limitations need to be acknowledged. First, incident OUD was identified solely through ICD-10 diagnosis codes, which may undercapture true OUD cases due to known under-coding, variability in diagnostic practices, and reluctance to document substance use disorders in clinical settings. As a result, some individuals with OUD may have been misclassified as noncases. Although prior work has similarly relied on diagnosis-based definitions [29,34-36], broader definitions incorporating buprenorphine treatment or combinations of high-risk prescribing behavior indicators may improve case ascertainment. Second, unmeasured predictors, such as socioeconomic determinants and illicit opioid use, could influence risk trajectories. Third, while the study uniquely assessed OUD risk within 3 months of opioid initiation, its reliance on older EHR data may limit real-time applicability due to potential delays in data availability and processing.

From a clinical use perspective, the model’s ability to stratify patients by risk supports early intervention and more efficient resource allocation in primary care settings. Our analysis confirms that the model’s risk estimates are not only accurate in distinguishing higher- from lower-risk patients but also well calibrated after post hoc adjustment. Decision-curve analysis further demonstrated that the model offers meaningful clinical utility across thresholds where default “treat-all” strategies offer little value in the context of a low outcome incidence. At the top decile threshold, the model achieved a positive net benefit equivalent to identifying approximately 2.6 additional true OUD cases per 1000 patients—representing about 50% of the maximum achievable use given the cohort’s baseline incidence. These findings align with established decision-analytic frameworks [70] and highlight the model’s potential to effectively prioritize individuals most likely to benefit from preventive intervention. Integrating such models into clinical decision support (CDS) tools could enable providers to proactively flag and manage patients at elevated risk for OUD before symptoms escalate. However, ethical considerations must also be addressed, including the potential for unintentionally reinforcing existing health disparities. Although the overall FNR and FPR parity ratio was within acceptable bounds, Black patients exhibited higher false negative rates across percentiles, suggesting that true OUD cases may be underidentified in this group. To address this disparity, we outline potential mitigation strategies for future model refinement. These include group-specific threshold adjustment to harmonize error rates, reweighting or cost-sensitive learning to shift model attention toward underidentified subgroups, and postprocessing approaches such as calibration-by-group or equalized-odds adjustment. Importantly, any fairness intervention should balance reducing disparities with avoiding unintended undertreatment or new forms of inequity. These considerations underscore the need for iterative fairness evaluation as part of a responsible deployment framework.

To support real-world deployment, our translational roadmap includes a structured, multiphase implementation strategy grounded in principles of human-centered design and clinical workflow alignment. In the retrospective phase, we will operationalize the OUD risk model within the UF Health Integrated Data Repository, a comprehensive enterprise clinical data warehouse. This step will enable validation using real-time patient data and ensure backend compatibility with the Epic EHR system. The model will generate individual-level risk scores for patients with recent opioid prescriptions, using secure internal infrastructure and consistent EHR data formatting. In the next phase, we will begin with a feasibility test and pilot integration of the model into the EHR system (Epic) within UF Health, gathering real-world feedback from clinicians before scaling up. Notably, UF Health’s Epic EHR infrastructure has already implemented a system-wide approach for an artificial intelligence–driven CDS tool that predicts patients with a high risk of opioid overdose in primary care workflows [71,72].

In the near term, we plan to deploy the OUD risk prediction model through EHR-based CDS tools that seamlessly integrate into routine care. The model’s risk score will trigger automated alerts during opioid prescribing encounters, prompting timely interventions for those identified as high risk. For example, in a primary care clinic with approximately 1000 patients initiating opioid therapy annually, applying a top decile (10%) risk threshold would flag approximately 100 individuals as high risk, of whom ~4 would be expected to develop OUD within 3 months. Although the PPV remains low due to the rarity of incident OUD, the alert volume is manageable in routine practice, particularly when recommended interventions are supportive and low-intensity (eg, increased follow-up, medication review, and motivational interviewing). The remaining ~90% low-risk patients would have an extremely high NPV, allowing providers to continue appropriate pain management for those individuals without unnecessary alarm. To further address alert fatigue, we propose specific operational strategies such as (1) embedding alerts within existing opioid or pain management workflows (eg, triggering only when signing a new medication order) to minimize disruption, (2) using tiered alerting triggered by repeated high-risk predictions, and (3) combining risk scores with other clinically meaningful indicators (eg, concurrent benzodiazepine prescribing) to enhance alert specificity. To minimize potential harms such as stigma and inappropriate opioid restriction, alerts should be framed using universal-precaution language that encourages supportive monitoring rather than implying misuse.

Although our findings indicate that the model has characteristics that could support future integration into clinical workflows, real-world feasibility and impact have not yet been evaluated. Additional implementation studies, including usability testing, workflow integration, and prospective evaluation, are needed before broad deployment in routine practice. While our separate pilot trial, Developing and Evaluating a Machine Learning Opioid Prediction & Risk-Stratification E-Platform, is currently assessing the feasibility of integrating an EHR-based overdose alert into clinical workflows within UF Health primary care clinics [71,72], analogous testing would be required for the model developed in this study.

This study demonstrates that an ML algorithm derived from the OneFlorida+ network can accurately predict incident OUD and effectively stratify risk across diverse populations and health systems, as evidenced by successful external validation in UPMC. The next step is to prospectively validate this algorithm within the UF Health clinical workflow to leverage existing decision support infrastructure and facilitate early intervention. Future research should expand the model to incorporate unstructured clinical notes, prescription drug monitoring program data, and social drivers of health (eg, housing instability, employment status, and incarceration history) to further improve prediction accuracy and equity.